Immunostimulants versus placebo for preventing exacerbations in adults with chronic bronchitis or chronic obstructive pulmonary disease
Abstract
Background
Individuals with chronic obstructive pulmonary disease (COPD) or chronic bronchitis may experience recurrent exacerbations, which negatively impact prognosis and quality of life, and can impose a significant socioeconomic burden on the individual and wider society. Immunostimulants are a broad category of therapies that may theoretically enhance non‐specific immunity against several respiratory insults, thereby reducing exacerbation risk and severity. However, evidence to date for their use in this population is limited.
Objectives
To determine the efficacy of immunostimulants in preventing respiratory exacerbations in adults with chronic obstructive pulmonary disease, chronic bronchitis, or both.
Search methods
We used standard, extensive Cochrane search methods. The latest literature search was conducted on 25 January 2022.
Selection criteria
We included parallel randomised controlled trials (RCTs) that compared immunostimulant therapy, administered by any method and with the intention of preventing (rather than treating) exacerbations, with placebo for a minimum treatment duration of one month in adults with chronic bronchitis or COPD, or both. We excluded participants with other respiratory conditions.
Data collection and analysis
We used standard Cochrane methods. Our primary outcomes were number of participants with no exacerbations during the study period and all‐cause mortality, secondary outcomes were respiratory‐related mortality, quality of life, number of participants requiring antibiotics, exacerbation duration, respiratory‐related hospitalisation duration and adverse events/side effects. We used GRADE to assess certainty of evidence for each outcome.
Main results
This review included 36 studies involving 6192 participants. Studies were published between 1981 and 2015. Duration ranged from three to 14 months. The mean age of study participants varied between 35.2 and 82 years. Twelve studies examined participants with COPD only. Seventeen studies reported baseline lung function values; most indicated a moderate‐to‐severe degree of airflow limitation. Nineteen studies indicated inclusion of participants with a mean baseline exacerbation frequency of two or more in the preceding year. Immunostimulants investigated were OM‐85, AM3, RU41740 (Biostim), Ismigen, Diribiotine CK, thymomodulin, pidotimod, D53 (Ribomunyl), Lantigen B, Symbioflor, and hyaluronan; routes of administration were oral, sublingual, and subcutaneous. The risk of bias of the included studies was mostly low or unclear.
Participants receiving immunostimulants for a mean duration of six months were slightly more likely to be free of exacerbations during that time (odds ratio (OR) 1.48, 95% confidence interval (CI) 1.15 to 1.90; 15 RCTs, 2961 participants; moderate‐certainty evidence). The overall number needed to treat with immunostimulants for a mean of six months, to prevent one participant from experiencing an exacerbation, was 11 (95% CI 7 to 29). This outcome was associated with a moderate degree of unexplained heterogeneity (I2 = 53%). Type of immunostimulant, baseline lung function, baseline exacerbation frequency, treatment duration, and follow‐up duration did not modify the effect size, although due to heterogeneity and limited study and participant numbers within some subgroups, the validity of the subgroup treatment effect estimates were uncertain.
Immunostimulants probably result in little to no difference in all‐cause mortality (OR 0.64, 95% CI 0.37 to 1.10; 5 RCTs, 1558 participants; moderate‐certainty evidence) and respiratory‐related mortality (OR 0.40, 95% CI 0.15 to 1.07; 2 RCTs, 735 participants; low‐certainty evidence) compared to placebo; however, the effects were imprecise and data quality limited the certainty of these results.
There was a small improvement in health‐related quality of life, as measured by the St George's Respiratory Questionnaire (SGRQ), with immunostimulant compared to placebo (mean difference −4.59, 95% CI −7.59 to −1.59; 2 RCTs, 617 participants; very‐low certainty evidence). The effect estimate just met the minimum clinically important difference (MCID) score of 4 units; however, the CI width means the possibility of a non‐meaningful difference cannot be excluded.
The pooled result from five studies indicated that immunostimulants likely reduce the number of participants requiring antibiotics over a mean duration of six months (OR 0.34, 95% CI 0.18 to 0.63; 542 participants; moderate‐certainty evidence). This outcome had a low‐to‐moderate degree of heterogeneity (I2 = 38%), but the direction of effect was consistent across all studies.
There was no evidence of a difference in the odds of experiencing an adverse event with immunostimulant compared to placebo, over a mean duration of six months (OR 1.01, 95% CI 0.84 to 1.21; 20 RCTs, 3780 participants; high‐certainty evidence). The CI limits for the associated risk ratio (RR) did not cross thresholds for appreciable harm or benefit (RR 1.02, 95% CI 0.92 to 1.13). An additional seven studies reported no events rates in either study arm.
Meta‐analyses were not performed for the outcomes of exacerbation duration and respiratory‐related hospitalisation duration, due to high levels of heterogeneity across the included studies (exacerbation duration: I2 = 92%; respiratory‐related hospitalisation duration: I2 = 83%). Results from an effect direction plot and binomial probability test for exacerbation duration indicated that a significant proportion of studies (94% (95% CI 73% to 99%); P = 0.0002) favoured intervention, possibly indicating that immunostimulants are efficacious in reducing the mean exacerbation duration compared to placebo. However, the degree of uncertainty associated with this estimate remained high due to data quality and heterogeneity. Three studies reported mean duration of respiratory‐related hospitalisation, two of which demonstrated a direction of effect that favoured immunostimulant over placebo.
Authors' conclusions
In participants with chronic bronchitis or COPD, we are moderately confident that treatment with immunostimulants is associated with a small reduction in the likelihood of having an exacerbation and a moderate reduction in the requirement for antibiotics. Low numbers of events limit interpretation of the effect of immunostimulants on all‐cause and respiratory‐related mortality. We are uncertain whether immunostimulants improve quality of life, and whether they are associated with a reduction in exacerbation and respiratory‐related hospitalisation durations, although immunostimulants were generally associated with a positive effect direction in the studies that examined these outcomes. Immunostimulants appear to be safe and well‐tolerated, and are not associated with an increased risk of adverse events.
PICOs
Plain language summary
Is taking an immunostimulant on top of standard medications beneficial for people with chronic obstructive pulmonary disease, chronic bronchitis, or both?
Key messages
1. In people with chronic bronchitis or COPD, immunostimulants probably reduce the likelihood of a person having an exacerbation and of requiring antibiotics for an exacerbation.
2. We are uncertain about the effect of immunostimulants on reducing the risk of death, improving quality of life, or on reducing the duration of flare‐ups or hospital stays.
3. Immunostimulants are not associated with an increased risk of side effects.
What are chronic obstructive pulmonary disease and chronic bronchitis?
Chronic obstructive pulmonary disease (COPD) and chronic bronchitis are common conditions that permanently affect the airways of the lungs. They are mainly caused by exposure to cigarette smoke or other air pollutants. People with COPD or chronic bronchitis may develop persistent symptoms of breathlessness, cough, and phlegm production, and are prone to flare‐ups (exacerbations) of these symptoms. Flare‐ups can be debilitating, worsen lung function over time, and cause further exacerbations. Several standard treatments exist to help prevent flare‐ups, which are recommended in almost all people who have a diagnosis of COPD. These mainly include quitting smoking, participating in exercise programmes, obtaining vaccinations to prevent infection, and using specific medications through an inhaler device.
What are immunostimulants?
Immunostimulants are a type of medication that are not widely used for the long‐term management of COPD or chronic bronchitis. Some scientists and doctors have suggested that immunostimulants, added to standard treatment, may help reduce the frequency and severity of flare‐ups in this patient group, by boosting the immune system response to triggers for exacerbations (such as infection with viruses or bacteria).
What did we want to find out?
We wanted to find out whether giving an immunostimulant medication, on top of standard treatment, reduced the frequency of flare‐ups in adults with COPD or chronic bronchitis. We also wanted to know whether immunostimulants reduced the risk of death, improved quality of life, and reduced the duration and severity of flare‐ups.
What did we do?
We searched for studies involving adults with COPD or chronic bronchitis or both, that took place over at least 12 weeks. Studies must have randomly divided participants into receiving either an immunostimulant or placebo (inactive replacement for a medicine), and directly compared the two groups. We included immunostimulants of any type and route of administration, although we did have prespecified criteria as to what constituted an 'immunostimulant' due to this term being relatively broad.
What did we find?
We found 36 studies involving 6192 participants. These looked at a variety of immunostimulants over durations that ranged from three to 14 months.
Results showed that participants receiving immunostimulants were slightly more likely to be free of exacerbations over an average duration of six months, compared to those who had received placebo. Treating 11 people with immunostimulants for six months would prevent one person from experiencing a flare‐up. We also found that immunostimulants likely reduced the number of participants requiring antibiotics for a flare‐up. Immunostimulants appeared to be safe, well‐tolerated, and not associated with an increased risk of side effects.
The impact of immunostimulants on death, quality of life, duration of flare‐ups, and the duration of hospital stays due to a flare‐up was unclear. Immunostimulants were generally favoured over placebo in most studies that looked at these outcomes.
What were the limitations of the evidence?
Overall, we are moderately confident in these results. Our confidence was reduced by how different some results looked between studies, the small number of studies included in some analyses, and in some instances by not having enough data or participant numbers to determine whether immunostimulants were truly better or the same as placebo.
How up‐to‐date is this evidence?
The evidence is up‐to‐date to January 2022.
Authors' conclusions
Summary of findings
| Patient or population: adults with chronic bronchitis or chronic obstructive pulmonary disease | ||||||
| Outcomes | Anticipated absolute effects* (95% CI) | Relative effect | № of participants | Certainty of the evidence | Comments | |
|---|---|---|---|---|---|---|
| Risk with placebo | Risk with immunostimulant | |||||
| Number of participants with no exacerbations during the study period | Low | OR 1.48 | 2961 | ⊕⊕⊕⊝ | Immunostimulants likely result in a slight increase in the number of participants with no exacerbations (or inversely, result in a reduction in the number of participants with ≥ 1 exacerbations). | |
| 5 per 100a | 7 per 100 | |||||
| High | ||||||
| 68 per 100a | 76 per 100 | |||||
| Mortality (all‐cause) | Low | OR 0.64 | 1558 | ⊕⊕⊕⊝ | Immunostimulants probably result in little to no difference in all‐cause mortality. | |
| 21 per 1000a | 14 per 1000 | |||||
| High | ||||||
| 58 per 1000a | 38 per 1000 | |||||
| Mortality (respiratory‐related) | 4 per 100 | 2 per 100 | OR 0.40 | 735 | ⊕⊕⊝⊝ | Immunostimulants may result in little to no difference in respiratory‐related mortality. |
| Quality of life | The mean quality of life was 37.5 points | MD 4.59 points lower | ‐ | 617 | ⊕⊝⊝⊝ | Immunostimulants may be associated with improvement in health‐related quality‐of‐life scores, but the evidence is very uncertain. |
| Number of participants requiring antibiotics | Low | OR 0.34 | 542 | ⊕⊕⊕⊝ | Immunostimulants likely result in a reduction in the number of participants requiring antibiotics. | |
| 60 per 100a | 34 per 100 | |||||
| High | ||||||
| 73 per 100a | 48 per 100 | |||||
| Adverse events/side effects | Low | OR 1.01 | 3780 | ⊕⊕⊕⊕ | Immunostimulants do not increase the number of participants with an adverse event. | |
| 6 per 100a | 6 per 100 | |||||
| High | ||||||
| 44 per 100a | 44 per 100 | |||||
| *The risk in the intervention group (and its 95% confidence interval) is based on the assumed risk in the comparison group and the relative effect of the intervention (and its 95% CI). | ||||||
| GRADE Working Group grades of evidence | ||||||
| See interactive version of this table: https://gdt.gradepro.org/presentations/#/isof/isof_question_revman_web_429628585692116898. | ||||||
| a The lowest and highest risk values are the second‐lowest and second‐highest proportions of participants with no exacerbations in the control groups from the studies included in this review. | ||||||
Background
Description of the condition
Chronic obstructive pulmonary disease (COPD) is defined by the Global Initiative for Chronic Obstructive Lung Disease (GOLD) as "a common, preventable and treatable disease which is characterised by persistent respiratory symptoms and airflow limitation that is due to airway and/or alveolar abnormalities, usually caused by significant exposure to noxious particles or gases" (GOLD 2022). Globally, COPD is a major cause of morbidity and mortality. In 2015, there was an estimated prevalence of 174 million cases, with three million deaths attributable to COPD (GBD 2015). Currently, COPD is considered the third leading cause of death worldwide (WHO GHE 2016). These numbers are projected to increase further over the next 30 years due to a combination of population ageing and ongoing exposure to COPD risk factors (Lopez 2006; WHO 2018). The economic burden associated with COPD is also substantial, with direct and indirect costs placing significant financial strain on individuals, their families, wider society, and healthcare systems worldwide (ATS Foundation 2014; Jinjuvadia 2017).
The symptoms of COPD include dyspnoea (breathlessness), chronic cough, and sputum production. COPD encompasses a range of clinical phenotypes, including emphysema and chronic bronchitis, with chronic bronchitis classically being defined as chronic cough and sputum production for at least three months per year for two consecutive years (Ferris 1978). Alternative definitions of chronic bronchitis exist, including cough and phlegm almost every day or several times a week (Kim 2015). Whilst chronic bronchitis is not technically defined by airflow limitation, it may precede the development of this, and is still thought to be associated with airway disease and inflammation, an increased risk in the total number and severity of respiratory exacerbations, and functional limitations (Kim 2011; Woodruff 2016).
A COPD exacerbation is defined as an acute worsening of respiratory symptoms that results in additional treatment (Wedzicha 2007). Exacerbations are often associated with increased airway inflammation, gas trapping, and mucous production (GOLD 2022); these changes typically lead to symptoms of increased dyspnoea, alteration in sputum colour or volume, increased cough and wheeze, or a combination of these. Most COPD exacerbations are triggered by viral or bacterial respiratory infections (or both); however, environmental changes and air pollution may also play a role in either causing or worsening exacerbations (GOLD 2022; Woodhead 2011). Studies have suggested that viruses are the causative pathogen in 34% to 56% of COPD exacerbations (Mohan 2010; Papi 2006; Rohde 2003), with bacterial infections reportedly associated with up to 50% of exacerbations (Papi 2006). Additionally, viral and bacterial coinfection is common, and has been shown to correlate with an increased severity of exacerbations and longer duration of hospitalisation (Papi 2006; Singanayagam 2012).
It is widely known that respiratory exacerbations in COPD are associated with increased mortality, accelerated decline in lung function, increased hospitalisation and readmission rates, and decreased quality of life (Kanner 2001; Soler‐Cataluña 2005). In addition, a history of previous exacerbations is said to be the single biggest risk factor for future exacerbations (Hurst 2010). Some people with COPD are more prone to frequent exacerbations (defined as two or more exacerbations per year) and this group has been shown to have worse outcomes and morbidity than those who experience less‐frequent exacerbations (Seemungal 1998). Aside from impacting the health status and prognosis of individuals, exacerbations also impose a significant socioeconomic burden on society, particularly those that necessitate hospital admission.
A number of evidence‐based therapies exist to reduce symptoms and exacerbations, and improve lung function, exercise tolerance, and quality of life, in people with COPD. Key aspects of COPD management include smoking cessation, exercise, pulmonary rehabilitation, and regular vaccinations for both influenza and pneumococcal infections (GOLD 2022). Other non‐pharmacological options for some people include treatment of hypoxaemia with long‐term oxygen therapy (Cranston 2005), treatment of hypercapnia with long‐term non‐invasive ventilation (Köhnlein 2014), and surgical or bronchoscopic lung volume reduction procedures (Marruchella 2018). Pharmacologically, the mainstay of treatment in stable COPD involves inhaled bronchodilators, including beta‐agonists and anti‐muscarinic agents (GOLD 2022; Kew 2013; Tashkin 2008). If people still have a high symptom or exacerbation burden, the addition of inhaled corticosteroids to a long‐acting beta‐agonist is recommended (Nannini 2012). A number of oral anti‐inflammatory agents also reduce exacerbations in COPD, and are currently recommended for use in some people, including phosphodiesterase‐4 inhibitors, mucolytic agents, and macrolide antibiotics (Chong 2013; Ni 2015; Poole 2015).
Description of the intervention
A glossary of the main immunological terms used is provided in Appendix 1.
The 2022 GOLD guidelines have specifically mentioned use of immunostimulant or immunoregulatory agents for preventing exacerbations in people with COPD (GOLD 2022). Immunostimulants are defined as agents that create a state of non‐specific immunity and enhance the immune response towards infection or malignancy (Hadden 1993). They have existed for many years, and have long been suggested to have efficacy in preventing or reducing the severity of acute respiratory tract infections (ARTIs).
In some countries, immunostimulants are regularly used for the prevention of ARTIs in children, and for reducing the frequency and severity of exacerbations in adults with COPD or chronic bronchitis (Del‐Rio‐Navarro 2007). However, their widespread and routine use has been limited due to a shortage of high‐quality data regarding their efficacy, and a lack of understanding of their mechanisms of action and long‐term safety profiles. The 2022 GOLD guidelines acknowledge that, whilst older studies have demonstrated the efficacy of immunostimulants in reducing the severity and frequency of COPD exacerbations, further studies are needed to examine their effects in people who are receiving current 'gold‐standard' COPD maintenance therapy (Collet 1997; Li 2004).
The immunostimulant agents that have been studied and used for the purpose of preventing ARTIs fall into three main categories: bacteria‐derived agents, synthetic agents, and thymic extracts (Del‐Rio‐Navarro 2007). Most immunostimulants used in the prevention of ARTIs are orally or sublingually administered bacteria‐derived agents. These can be further categorised into those that are inactivated whole‐cell formulations, those that contain a mixture of antigenic fragments derived from several bacterial strains (bacterial lysates), and those that consist of a specific immunogenic component of a bacterium, such as ribosomal fractions or glycoproteins (Cazzola 2008; Del‐Rio‐Navarro 2007; Giovannini 2014).
Bacterial lysates are composed of constituted fragments of bacterial antigens, obtained through the chemical or mechanical lysis (breakdown) of multiple inactivated bacterial strains that are commonly associated with respiratory infections, such as Staphylococcus aureus, Streptococcus pneumoniae, Haemophilus influenzae, Klebsiella pneumoniae, Moraxella catarrhalis, Streptococcus pyogenes, and Streptococcus viridans (De Benedetto 2013).
The commercial availability of these agents varies by country, region, and clinical indication for use. As examples, a range of bacterial lysates have been approved for use for the prevention of recurrent respiratory tract infections in several European countries, including OM‐85, Ismigen, Ribomunyl, Lantigen B, and LW50020 (EMA 2019). OM‐85 is available and used across Central and South America for the treatment and prophylaxis of respiratory infections in children and adults (Pivniouk 2022). In China, several immunostimulant agents are approved for use including OM‐85 and pidotimod (a synthetic immunostimulant agent) (CMDD 2016; Pivniouk 2022). Conversely, immunostimulants are not currently approved for use in North America or Australasia.
How the intervention might work
Immunostimulants used for the purpose of preventing respiratory tract infections in people with COPD or chronic bronchitis aim to heighten the host immune response against infective insults that may subsequently trigger an exacerbation. However, despite there being much research and knowledge gained about the effects of individual immunostimulatory agents on the immune system, the exact mechanism of action of both synthetic and bacteria‐derived agents at a molecular level is not completely understood (De Benedetto 2013; Del‐Rio‐Navarro 2007).
OM‐85, a bacterial lysate derived by the chemical lysis of a number of the aforementioned bacterial strains, is thought to exert its effects through both cell‐mediated and humoral immune system pathways (De Benedetto 2013; Rozy 2008). These include augmentation of the T helper cell lymphocyte (Th1) response (Huber 2005), induction of specific immunoglobulin A antibody secretion by B lymphocyte cells (Rial 2004; Rossi 2003), direct activation of lung macrophages and monocytes (Mauel 1989; Rozy 2008), upregulation of adhesion molecules (Duchow 1992), and stimulation of phagocytic cell activity (Rozy 2008). It is thought that the immunostimulant components of OM‐85 bind to toll‐like receptors (TLRs), triggering signalling pathways that lead to activation and potentiation of the innate immune response (Alyanakian 2006; Huber 2005; Navarro 2011; Nikolova 2009).
Mechanical bacterial lysates, specifically polyvalent mechanical bacterial lysates (PMBLs), have been demonstrated to stimulate dendritic cell maturation (Morandi 2011), increase the number of circulating natural killer (NK) cells (Lanzilli 2013), increase specific immunoglobulin A antibody secretion by B cells, (Rossi 2003), and activate both memory B lymphocytes and regulatory T lymphocytes (Lanzilli 2013). Studies have shown that the degree of the immune response created by the administration of mechanical bacterial lysates to patients directly correlates with positive clinical outcomes, such as reduced exacerbation frequency (Braido 2011; Lanzilli 2006; Ricci 2014).
Other bacterial extracts, made up of bacterial proteins or ribosomal fragments, also have specific immunomodulatory effects. For example, the immunostimulant RU41740 (Biostim), made up of Klebsiella pneumoniae glycoproteins and membrane fragments, activates macrophages, stimulates the B lymphocyte cell response (Boissier 1988), and enhances antigen presentation (Pedraza‐Sánchez 2006). Like bacterial lysates, RU41740 is thought to initiate an immune response through binding of its molecular components (such as lipopolysaccharide) to TLRs (Miller 2005).
The immunostimulants containing thymic extracts also appear to interact with other TLRs and precursor T lymphocytes to increase dendritic cell, NK cell, and T cell activity, thus enhancing both innate and cell‐mediated immune responses (Del‐Rio‐Navarro 2007; Tuthill 2013). Synthetic immunostimulant compounds are reportedly better understood in terms of their molecular mechanism of action (Del‐Rio‐Navarro 2007). Examples of synthetic immunostimulants include: tucaresol, which acts by promoting the interaction between antigen‐presenting cells and T cells (Rhodes 1996); imiquimod, which acts through TLR7 and TLR8 (Spaner 2005); and pidotimod, which acts by enhancing cell‐mediated immunity (Benetti 1994).
A number of trials and systematic reviews have analysed the use of immunostimulant in preventing ARTIs in children and adults, and in preventing exacerbations in adults with chronic bronchitis or COPD. One previous Cochrane Review evaluated immunostimulants for preventing respiratory tract infections in children. The review authors included 34 placebo‐controlled trials, and reported that immunostimulants were associated with approximately 40% fewer acute respiratory infections compared with placebo. However, they also commented that "trial quality was generally poor and a high level of statistical heterogeneity was evident" (Del‐Rio‐Navarro 2012). One meta‐analysis examining the efficacy of PMBLs in preventing respiratory tract infections in children and adults looked at data across 15 randomised controlled trials. Treatment with PMBLs was associated with a significant reduction in respiratory infections compared with placebo (Cazzola 2012).
One systematic review found that the bacterial extracts OM‐85, LW‐50020, and SL‐04 were associated with improved symptoms in people with COPD, chronic bronchitis, or both, and the meta‐analysis suggested a lessened exacerbation duration. However, there was no evidence of a difference between the extracts and placebo in preventing exacerbations (Steurer‐Stey 2004). Another systematic review examined the efficacy of OM‐85 in preventing exacerbations in people with COPD or chronic bronchitis (or both). There was a non‐significant trend in favour of OM‐85; however, benefit was not clearly demonstrated across a range of important clinical outcomes (Sprenkle 2005). In 2015, one meta‐analysis and systematic review examined the effects of OM‐85 in people with COPD on exacerbation rate, in addition to several other minor clinical end points. OM‐85 was associated with a 20% reduction in exacerbation rate and 39% reduction in the incidence rate of people using antibiotics compared with placebo (Pan 2015). However, the authors concluded that there was insufficient evidence to support the routine use of OM‐85 in people with COPD, suggesting that further larger‐scale trials needed to be undertaken.
The cost‐effectiveness of the immunostimulant OM‐85 for preventing respiratory exacerbations in adults with COPD or chronic bronchitis (or both) has previously been examined. Prospective cost‐effectiveness and cost–benefit analyses were conducted alongside a randomised controlled trial comparing OM‐85 to placebo in adults with COPD (Collet 1997). Authors found that its use was associated with a reduction in the direct and indirect costs of a severe exacerbation warranting hospitalisation (Collet 2001). Other studies have also found that OM‐85 appears to be cost‐effective for the prevention of acute respiratory exacerbations in adults with chronic bronchitis (Bergemann 1994; Xuan 2014).
Why it is important to do this review
Immunostimulant agents in COPD or chronic bronchitis could theoretically enhance non‐specific immunity against a number of respiratory insults, which is an important concept given the number of pathogens, including myriad viruses, that can precipitate an exacerbation. However, the use of immunostimulants in this population thus far has been controversial. This is largely due to concerns about quantity and quality of evidence in the past, a lack of understanding of their long‐term effects, and existing uncertainty around their exact mechanisms of action. As mentioned, GOLD guidelines recognise that some older studies have reported a decrease in the severity and frequency of COPD exacerbations, but more trials are needed in people receiving currently recommended maintenance therapy (GOLD 2022). Overall, there have been mixed results from previous trials and systematic reviews regarding clinical efficacy of immunostimulants in adults with COPD or chronic bronchitis; many have reported positive impacts on exacerbation rates (or a trend towards this) and other clinically relevant outcomes, but have been unable to suggest an overall benefit in these patients due to limited availability of data, poor trial quality, or both.
This systematic review aims to critically appraise all available data regarding the efficacy and use of immunostimulants as a preventive therapy in adults with stable COPD or chronic bronchitis (or both). This may help to further understand their clinical value and safety, and may highlight areas requiring further research and development.
Objectives
To determine the efficacy of immunostimulants in preventing respiratory exacerbations in adults with chronic obstructive pulmonary disease, chronic bronchitis, or both.
Methods
Criteria for considering studies for this review
Types of studies
We included parallel randomised controlled trials (RCTs) comparing immunostimulant therapy, administered by any route, with placebo. We included studies reported in full text, those published as an abstract only, and unpublished data. We excluded cross‐over trials due to the nature of COPD as a progressive disease and due to the possibility of a carry‐over effect from the first treatment period. However, if these studies presented separate comparison data prior to any cross‐over occurring then this data set was considered for review inclusion.
Types of participants
We included studies of adults (older than 18 years of age) with a diagnosis of COPD (defined by a postbronchodilator forced expiratory volume in one second (FEV1)/forced vital capacity (FVC) ratio less than 0.7) or chronic bronchitis (defined by either the classic definition of chronic cough and sputum production for at least three months per year for two consecutive years (Ferris 1978), or alternative definitions such as cough and phlegm almost every day or several times a week (Kim 2015)), or both COPD and chronic bronchitis. An underlying principle was that study participants met well‐established criteria at the time, whether for COPD or chronic bronchitis.
We excluded studies of participants with asthma, bronchiectasis, or genetic/other lung conditions that predispose or lead to chronic airflow obstruction, such as cystic fibrosis, or people with known specific immunodeficiencies. However, if a study included these types of participants and additionally included people with COPD or chronic bronchitis (or both), we analysed data for these subsets of participants if presented separately.
Types of interventions
Participants must have received immunostimulant therapy or placebo, administered by any route (oral, sublingual, subcutaneous, or intravenous), for at least one month. Where a study involved an intermittent dosing regimen, in which the cumulative treatment days totalled less than one month, but the overall treatment duration was at least one month, we elected to include this study in the review.
We excluded studies of specific immunostimulants, such as influenza or pneumococcal vaccines, immunotherapy used to treat cancer (either directly, or the immune deficiency resulting from chemotherapy), allergic disease, and treatment to replace immunoglobulins in known specific immune deficiency disorders. We excluded trials referring to vitamins, nutritional supplements, herbal extracts, or homeopathic remedies. We excluded studies that focussed on the treatment of acute exacerbations with immunostimulants (as opposed to prevention and prophylaxis). We did not consider studies that focussed on improvement in immunological parameters as the sole outcome.
Types of outcome measures
We only included studies where the primary outcomes were measured for at least 12 weeks.
Reporting of one or more of the listed outcomes was not an inclusion criterion for the review.
Exacerbations were chosen as the primary outcome as immunostimulant therapy is most commonly administered as a preventive therapy, and thus far exacerbations have represented the main clinically relevant end point used for efficacy of immunostimulant agents and many other anti‐inflammatory drugs used in COPD or chronic bronchitis (GOLD 2022). As mentioned, exacerbations significantly affect patient morbidity and mortality and have a significant impact on overall disease burden; therefore, we considered that exacerbations and mortality were important as primary outcomes in this review.
Primary outcomes
-
Number of participants with no exacerbations during the study period.
-
Mortality (all‐cause).
There was considerable variability in the degree of reporting and description of what constituted a respiratory exacerbation between included studies; hence, we accepted the study authors' definition of an exacerbation.
As outlined in our review protocol (Fraser 2019), we had intended to look at and delineate the number of participants with exacerbations of COPD or chronic bronchitis (or both) that would be considered moderate or severe in accordance with GOLD criteria definitions (GOLD 2022), and use this information to inform our primary outcomes along with all‐cause mortality. However, following the data extraction process and prior to performing any statistical analysis, it was apparent that these outcomes could not be determined from the information and data presented in the included studies.
In order to allow us to combine and interpret our review outcomes, a deviation from our original intended primary outcomes was considered necessary. Therefore, we chose the dichotomous outcome of 'number of participants with no exacerbations during the study period.'
Secondary outcomes
-
Mortality (respiratory‐related).
-
Quality of life (participant‐reported, measured by a validated scale, such as the St George's Respiratory Questionnaire (SGRQ) (Jones 1992) or Chronic Respiratory Diseases Questionnaire (CRQ) (Guyatt 1987)).
-
Number of participants requiring antibiotics.
-
Exacerbation duration.
-
Hospitalisation duration (respiratory‐related).
-
Adverse events/side effects.
The secondary outcomes are other important measures of the efficacy and safety of immunostimulant agents.
Our protocol outlined our intentions to include the 'total number of exacerbations' as a secondary outcome; however, including this was no longer necessary given the reformulation of the primary outcomes that would capture this information. Analysing exacerbation duration was considered to provide indirect information regarding the potential socioeconomic and quality‐of‐life impacts of the intervention. Quality‐of‐life assessments highlight information regarding the impact of the intervention on the objective and subjective wellbeing of the patient, and are an important outcome in any chronic disease. Given that little is understood about the long‐term safety of these agents and concerns around this have previously limited recommendation for widespread use, adverse events were included as a secondary outcome for this review.
We added 'Number of participants requiring antibiotics' and 'hospitalisation duration (respiratory‐related)' as secondary outcome measures following data extraction and prior to data synthesis occurring. These were metrics reported in several included studies, and were considered surrogate markers of exacerbation severity in lieu of being able to obtain clear data around how many of the exacerbations could be considered 'moderate' or 'severe' (or both) by GOLD definitions (GOLD 2022). Several studies reported 'Antibiotic duration'; however, we considered that a dichotomous outcome regarding antibiotic use would be more statistically robust. Conversely, the 'number of participants requiring hospitalisation' was rarely reported, and inclusion as an outcome would have contributed little to overall comparison.
Search methods for identification of studies
Electronic searches
We identified studies from searches of the following databases and trial registries:
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Cochrane Airways Trials Register (airways.cochrane.org/trials-register), via the Cochrane Register of Studies, all years to 25 January 2022;
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Cochrane Central Register of Controlled Trials (CENTRAL), via the Cochrane Register of Studies, all years to 25 January 2022;
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MEDLINE (OvidSP) ALL 1946 to 25 January 2022;
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Embase (OvidSP) 1974 to 25 January 2022;
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US National Institutes of Health Ongoing Trials Register ClinicalTrials.gov (www.clinicaltrials.gov);
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World Health Organization International Clinical Trials Registry Platform (apps.who.int/trialsearch).
The database search strategies are detailed in Appendix 2. The Cochrane Airways Information Specialist in collaboration with the review authors wrote the search strategies and executed the searches.
We searched all databases and trials registries from their inception to 25 January 2022, with no restrictions on language or type of publication. We handsearched conference abstracts and grey literature identified through the Cochrane Airways Trials Register and the CENTRAL database.
Searching other resources
We checked the reference lists of all primary studies and review articles for additional references. We searched relevant manufacturers' websites for study information.
Data collection and analysis
Selection of studies
We used Cochrane's Screen4Me workflow to help assess the search results. Screen4Me comprises three components: known assessments – a service that matches records in the search results to records that have already been screened in Cochrane Crowd and been labelled as an RCT or as Not an RCT; the RCT classifier – a machine learning model that distinguishes RCTs from non‐RCTs, and if appropriate, Cochrane Crowd – Cochrane's citizen science platform where the Crowd help to identify and describe health evidence (Thomas 2020). We used the first two components.
Following this initial assessment, two review authors (AF and PP) independently screened the titles and abstracts of the search results and coded them as 'retrieve' (eligible or potentially eligible/unclear) or 'do not retrieve.' We retrieved the full‐text study reports of all potentially eligible studies and two review authors (AF and PP) independently screened them for inclusion, recording the reasons for exclusion of ineligible studies. Where we could not obtain the full‐text study reports, we marked these studies as 'awaiting classification.' We resolved any disagreements through discussion. We identified and excluded duplicates and collated multiple reports of the same study so that each study, rather than each report, was the unit of interest in the review. Seventeen translators reviewed papers published in languages other than English; if studies met initial screening criteria and were subsequently marked for inclusion, the translators completed formal data extraction sheets for individual studies. We recorded the selection process in sufficient detail to complete a PRISMA flow diagram (Moher 2009), and documented justification for each study exclusion (see Characteristics of excluded studies table and Excluded studies).
Data extraction and management
We extracted and collated data using the online software tool Covidence (Covidence), and exported online data extraction forms for individual studies to Excel worksheets. We extracted the following study characteristics from included studies using the Covidence data extraction process as a template.
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Methods: study design, total duration of study, number of study centres and location, study setting, date of study.
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Participants: number (n), mean age, age range, gender, severity of condition, diagnostic criteria, baseline lung function, smoking history, baseline exacerbation frequency, inclusion criteria and exclusion criteria.
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Interventions: intervention, comparison, dosing regimen, type of compound, route of administration, duration of therapy.
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Outcomes: primary and secondary outcomes specified and collected, and time points reported.
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Notes: funding for studies and notable conflicts of interest of trial authors.
Two review authors (AF and PP) independently extracted outcome data from included studies. These data were then cross‐checked between the two review authors and a consensus decision reached by discussion for each included study. We extracted data if it was deemed to be potentially relevant to protocol‐specified outcomes; the Characteristics of included studies table and Table 1 identify the reported study outcomes for which we extracted data. One review author (AF) transferred data into the Review Manager 5 (Review Manager 2014), and double‐checked data against the data presented within the original study reports. The second review author (PP) spot‐checked study characteristics for accuracy against the study report.
| Study ID | Total na | Study duration (weeks) | Mean age (years) | Participant type | Presence of acute exacerbation as an inclusion criteria | Intervention (total duration) | Category | Route | Relevant outcomes measured |
|---|---|---|---|---|---|---|---|---|---|
| 344 | 26 | 67.7 | COPD | No | AM3 1 g 3 times daily (6 months) | Candida utilis polysaccharide/protein compound | Oral | Participants with/without an exacerbation, number of exacerbations, SGRQ score, AEs | |
| 364 | 13 | 57.7 | COPD | No | AM3 3 g daily (3 months) | Candida utilis polysaccharide/protein compound | Oral | SGRQ score | |
| 110 | 26 | 62.9 | CB/COPD (data extracted limited to COPD patient subset) | No | RU41740 (Biostim) 2 mg daily for 8 consecutive days in first month; 1 mg daily for 8 consecutive days in second/third months (3 months) | Bacteria‐derived | Oral | Participants with/without an exacerbation, number of exacerbations, exacerbation duration, number of participants requiring antibiotics, AEs | |
| 181 | 17 | 62.3 | CB | No | Pidotimod 800 mg daily (2 months) | Synthetic agent | Oral | Participants with/without an exacerbation, AEs | |
| 184 | 52 | 69.2 | COPD (asthma patient subset excluded from analysis) | No | D53 (Ribomunyl) 4 sequences of 15 days of aerosol treatment separated by 1‐week intervals. Subcutaneous injections days 7 and 14 of first sequence and day 14 of following sequences (2.7 months) | Bacteria‐derived | Aerosol and subcutaneous | Mean number of exacerbations per participant, exacerbation duration, participants with no or a reduction in antibiotic therapy, AEs | |
| 172 | 26 | 60.5 | CB/COPD | No | RU41740 (Biostim) 2 mg daily or 8 mg daily (2 intervention groups) for 1 week, alternate weeks (3 months) | Bacteria‐derived | Oral | Participants with/without an exacerbation, exacerbation duration, duration of antibiotic therapy, AEs, mortality (all‐cause) | |
| 40 | 17 | 70.0 | CB/COPD | No | AM3 500 mg 3 times daily (4 months) | Candida utilis polysaccharide/protein compound | Oral | Participants with/without an exacerbation, mean number of exacerbations per participant, number of exacerbations, exacerbation duration, AEs | |
| 288 | 52 | 69.0 | COPD | No | Ismigen 50 mg daily for 10 consecutive days/month for 3 months, then 3 months without treatment, then repeat of the initial regimen (9 months) | Polyvalent mechanical bacterial lysate | Sublingual | Participants with/without an exacerbation, exacerbation rate, days to first exacerbation, hospitalisation days (respiratory and all‐cause), participants requiring concomitant medications, QoL scale scores, AEs, mortality (all‐cause) | |
| 40 | 13 | 65.0 | COPD | No | AM3 500 mg 3 times daily (3 months) | Candida utilis polysaccharide/protein compound | Oral | Mean number of exacerbations per participant, AEs | |
| 236 | 13 | 64.9 | COPD plus "cell‐mediated immune deficiency" | No | Thymomodulin 60 mg twice daily (3 months) | Thymic extract | Oral | Exacerbation rate, QoL scale scores, AEs | |
| 178 | 13 | 66.5 | COPD | No | Ismigen 50 mg daily for 10 consecutive days/month for 3 months (3 months) | Polyvalent mechanical bacterial lysate | Sublingual | Exacerbation rate, exacerbation duration, duration of antibiotic therapy, hospitalisation rate (respiratory), hospitalisation duration (respiratory), hospitalisation days (respiratory), AEs, mortality (all‐cause) | |
| 494 | 22 | 65.7 | CB/COPD | No | Pidotimod 800 mg daily (2 months) | Synthetic agent | Oral | Exacerbation rate, days to first exacerbation, exacerbation duration, duration of antibiotic therapy, AEs | |
| 381 | 26 | 66.1 | COPD | No | OM‐85 7 mg daily for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (4 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, mean number of exacerbations per participant, hospitalisation rate (respiratory and all‐cause), participants requiring hospitalisation (respiratory and all‐cause), hospitalisation duration (respiratory), hospitalisation days (respiratory and all‐cause), mean number of hospital days per participant (respiratory), change in SF‐36 scale scores, AEs, mortality (respiratory and all‐cause) | |
| 104 | 26 | 48.2 | CB/COPD | Yes | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Exacerbation duration, total exacerbation days, participants requiring antibiotics, participants requiring bronchodilator therapy, FEV1, AEs | |
| 265 | 26 | 81.8 | CB/COPD | Yes | OM‐85 7 mg daily for 10 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, number of exacerbations, participants requiring antibiotic therapy, AEs | |
| 60 | 17 | 62.7 | CB plus "borderline immune deficiency" | No | Lantigen B (2 intervention groups) 15 drops twice daily for 1 month, then 1 month without treatment, then a 15 days of initial regimen (2.5 months) | Bacterial lysate | Sublingual | Mean number of exacerbations per participant, exacerbation duration, participants requiring antibiotics, AEs | |
| 59 | 26 | 45.5 | CB | Yes | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Mean number of exacerbations per participant, exacerbation duration, duration of concomitant therapy, FEV1 | |
| 357 | 26 | Not reported | COPD | No | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, number of exacerbations, exacerbation duration, days to first exacerbation, duration of concomitant therapy, hospitalisation rate and duration (all‐cause), SGRQ scores, FEV1, AEs/serious AEs, mortality (all‐cause) | |
| 29 | 39 | 57.0 | CB/COPD | No | RU41740 (Biostim) 2 mg daily for 8 consecutive days in first month; 1 mg daily for 8 consecutive days in second/third months (3 months) | Bacteria‐derived | Oral | Participants with/without an exacerbation, mean number of exacerbations per participant, exacerbation duration, AEs | |
| 64 | 26 | 46.0 | CB | No | D53 (Ribomunyl) 1 tablet (dose not specified) daily for 4 consecutive days/week for 3 weeks, then 1 tablet daily for 4 consecutive days/month for 5 months (5.75 months) | Bacteria‐derived | Oral | Mean number of exacerbations per participant, exacerbation duration, participants requiring antibiotics, AEs | |
| 136 | 60 | 47.3 | CB | No | Symbioflor 30 drops 3 times daily (6 months) | Bacteria‐derived | Oral vs sublingual (liquid preparation) | Participants with/without an exacerbation, number of exacerbations, days to first exacerbation, participants requiring antibiotics, AEs | |
| 114 | 26 | 51.7 | CB/COPD | No | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Exacerbation duration, duration of antibiotic therapy | |
| 81 | 26 | 57.0 | CB/COPD | Yes | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Number of exacerbations, duration of antibiotic therapy, duration of corticosteroid therapy, hospitalisation rate (respiratory) | |
| 90 | 52 | 66.0 | CB/COPD | No | OM‐85 7 mg daily for 10 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral | Mean number of exacerbations per participant, exacerbation duration, duration of antibiotic therapy, AEs | |
| 44 | 26 | 47.7 | COPD | No | Diribiotine CK 10 mL daily for 20 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral (liquid preparation) | Participants with/without an exacerbation, mean number of exacerbations per participant, exacerbation duration, participants requiring antibiotics, AEs | |
| 79 | 26 | 55.1 | CB | Yes | OM‐85 7 mg daily for 10 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral | AEs | |
| 340 | 22 | Not reported | COPD | No | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Exacerbation rate, exacerbation duration, AEs | |
| 354 | 26 | 82.0 | CB/COPD | No | OM‐85 7 mg daily for 10 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, number of exacerbations, mean number of antibiotic courses, participants requiring bronchodilator therapy, participants requiring corticosteroids, AEs, mortality (respiratory and all‐cause) | |
| 19 | 17 | 52.8 | CB | No | OM‐85 7 mg daily for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (4 months) | Bacterial lysate | Oral | Mean number of exacerbations per participant, exacerbation duration, hospitalisation duration (respiratory), mean number of hospitalisations per participant (respiratory), AEs | |
| 88 | 52 | Not reported | CB/COPD | No | Thymomodulin 80 mg 3 times daily (3 months) | Thymic extract | Oral | Mean number of exacerbations per participant, exacerbation duration, hospitalisation duration (all‐cause), mean number of concomitant medication courses, duration of concomitant therapies, AEs | |
| 55 | 26 | 55.7 | CB | No | Diribiotine CK 10 mL daily for 20 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral (liquid preparation) | Mean number of exacerbations per participant, exacerbation duration, mean number of antibiotic courses, duration of antibiotic therapy, AEs | |
| 273 | 26 | 57.6 | CB/COPD | Yes | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, mean number of exacerbations per participant, number of exacerbations, participants requiring concomitant medication, AEs, mortality (all‐cause) | |
| 50 | 26 | 35.2 | CB | No | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Mean number of exacerbations per participant, exacerbation duration, duration of antibiotic therapy, duration of bronchodilator therapy, AEs | |
| 428 | 22 | 63.1 | CB/COPD | Yes | OM‐85 7 mg daily for 10 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, participants requiring concomitant medication, AEs | |
| 29 | 35 | 61.0 | CB/COPD | No | Hyaluronan 7.5 mg subcutaneously weekly (6 months) | Synthetic agent | Subcutaneous | Participants with/without an exacerbation, number of exacerbations, total exacerbation days, total antibiotic days, AEs | |
| 62 | 26 | 57.9 | COPD | No | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Number of exacerbations, exacerbation duration, participants requiring antibiotics, participants requiring bronchodilator therapy |
AE: adverse events; CB: chronic bronchitis; COPD: chronic obstructive pulmonary disease; QoL: quality of life; SF‐36: 36‐item Short Form Survey; SGRQ: St George's Respiratory Questionnaire.
aTotal n: number of participants for whom outcome data were available.
Assessment of risk of bias in included studies
Two review authors (AF and PP) independently assessed risk of bias for each study using the RoB 1 tool for RCTs according to criteria outlined in the Cochrane Handbook for Systematic Reviews of Interventions (Higgins 2011). We resolved any disagreements by discussion. We assessed risk of bias according to the following domains.
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Random sequence generation.
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Allocation concealment.
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Blinding of participants and personnel.
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Blinding of outcome assessment.
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Incomplete outcome data.
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Selective outcome reporting.
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Other bias.
We judged each potential source of bias as high, low, or unclear, and provided a quote where relevant from the study report together with a justification for our judgement. We summarised the risk of bias judgements across different studies for each of the domains listed. When assessing attrition bias, we used an approximate cut‐off of 20% dropout for 'high' risk; however, the type of analysis performed (e.g. intention‐to‐treat (ITT)), the balance between trial arms, and reasons given for dropout were also taken into account. When considering treatment effects, we considered the risk of bias for the studies that contributed to that outcome.
Assessment of bias in conducting the systematic review
Generally, we conducted the review in accordance with the published protocol (Fraser 2019). We reported and justified deviations from the protocol in the Differences between protocol and review section of the review.
Measures of treatment effect
We analysed dichotomous data as odds ratios (ORs) with a 95% confidence interval (CI), and continuous data as mean differences (MD) with a 95% CI. For dichotomous outcomes, we intended to calculate the number needed to treat for an additional beneficial outcome (NNTB) and number needed to treat for an additional harmful outcome (NNTH).
Meta‐analyses were undertaken only where this was meaningful; that is, if the treatments, participants, and the underlying clinical question were similar enough for pooling to make sense.
We planned to describe skewed data narratively (e.g. as medians and interquartile ranges for each group).
Where multiple trial arms were reported in a single study, we included only the relevant arms. All treatment arms are specified in the Included studies section and Table 1. If two comparisons (e.g. drug A versus placebo and drug B versus placebo) were combined in the same meta‐analysis, we combined the active arms.
Several studies reported outcomes measured over discrete time intervals. Where this occurred, and the outcome was measured over a period of less than 12 weeks, we excluded these data from a meta‐analysis. If the data were analysed over at least 12 weeks, we elected to include the data that had been measured from the commencement of the study in a meta‐analysis (i.e. from baseline, as opposed to from a later time point within the study). This earlier period was chosen as it was considered to best encapsulate participants who were truly randomised, with no likelihood of a carry‐over treatment effect from earlier in the study potentially affecting outcome measures. The results from later time intervals of the relevant studies, whilst not included in meta‐analyses, are discussed within the text of the review.
We used the difference between end point scores for studies that reported quality of life using a validated scale.
We used ITT or 'full analysis set' analyses where they were reported (i.e. those where data had been imputed for participants who were randomly assigned but did not complete the study), instead of completer or per‐protocol analyses.
Unit of analysis issues
For dichotomous outcomes, we used participants, rather than events, as the unit of analysis (e.g. the number of participants with an exacerbation, rather than the number of exacerbations per participant). We avoided treating count data as dichotomous data to avoid unit‐of‐analysis error in recurring events.
Dealing with missing data
When missing data were thought to introduce serious bias, this was taken into consideration in the GRADE rating for affected outcomes.
We recorded any assumptions made around the reason or nature of missing data within the review. We performed sensitivity analyses to assess how variable the results may have been to any assumptions made about missing data.
Assessment of heterogeneity
We used the I2 statistic to measure heterogeneity among the studies in each analysis. Where there was substantial heterogeneity, we explored possible causes by performing prespecified subgroup analyses. As per Chapter 10 of the Cochrane Handbook for Systematic Reviews of Interventions (Deeks 2022), we considered the following ranges for assessing heterogeneity.
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0% to 40%: might not be important.
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30% to 60%: may represent moderate heterogeneity.
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50% to 90%: may represent substantial heterogeneity.
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75% to 100%: may show considerable heterogeneity.
Assessment of reporting biases
When we were able to pool more than 10 studies, we created and examined a funnel plot to explore possible small‐study and publication biases.
Data synthesis
We used a random‐effects model. Sensitivity analyses were performed using a fixed‐effect model.
Subgroup analysis and investigation of heterogeneity
Subgroup analyses were performed based on measures of disease severity (baseline FEV1 and exacerbation frequency).
Most studies that reported the primary outcomes included orally delivered immunostimulant agents, with only two studies analysing sublingually and subcutaneously delivered immunostimulants. Therefore, we considered it less relevant to include 'mode of delivery' as a subgroup, as originally outlined in our protocol (Fraser 2019). We considered further differentiation of the immunostimulant agents based on dose and treatment regimen; however, there was much variability between studies limiting the ability to subgroup by these categories. Therefore, we only performed subgroup analyses based on the immunostimulant type.
Because of the variation seen in treatment course lengths and follow‐up periods across all studies, we used the point of longest follow‐up for assessment of outcomes overall, and added subgroup analyses following the data extraction phase to investigate for heterogeneity that may have been explained by treatment or overall study duration (or both).
We made these above alterations by consensus discussion following data extraction, but prior to any data synthesis and analysis occurring.
We performed the following subgroup analyses.
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Type of immunostimulant agent.
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Severity of COPD based on lung function testing: mild or moderate (defined by FEV1 50% or greater predicted) versus severe or very severe (FEV1 less than 50% predicted) (GOLD 2022).
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Mean baseline exacerbation rate two or more in the preceding year versus fewer than two in the preceding year or unspecified.
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Treatment duration (three months or less versus greater than three months).
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Study duration (three to less than six months versus six months to less than 12 months versus 12 months or greater).
We used the following outcomes in subgroup analyses.
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Number of participants with no exacerbations during the study period.
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Mortality (all‐cause).
We used the formal test for subgroup interactions in Review Manager 5 (Review Manager 2014).
Sensitivity analysis
We carried out the following sensitivity analyses, removing from the primary outcome analyses:
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trials judged in the risk of bias table at high risk of bias for any of the six domains;
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trials where decisions or assumptions had been made around missing data.
We compared the results from a fixed‐effect model with the random‐effects model.
Summary of findings and assessment of the certainty of the evidence
The summary of findings table contained the following outcomes.
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Number of participants with no exacerbations during the study period.
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Mortality (all‐cause).
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Mortality (respiratory‐related).
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Quality of life.
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Number of participants requiring antibiotics.
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Adverse events/side effects.
We used the five GRADE considerations (risk of bias, consistency of effect, imprecision, indirectness, and publication bias) to assess the certainty of the body of evidence as it related to the studies that contributed data for the prespecified outcomes. We used the methods and recommendations described in Chapter 14 of the Cochrane Handbook for Systematic Reviews of Interventions (Schünemann 2022), using GRADEpro GDT software (GRADEpro GDT). We outlined all decisions to downgrade the certainty of the evidence using footnotes and comments to aid in readers' understanding of the review.
Results
Description of studies
This review is based on a published protocol (Fraser 2019).
Results of the search
For details of the search history, see Appendix 2, and for the PRISMA study flow diagram, see Figure 1.
Following deduplication, the database and trial registry searches run on 24 May 2019 and 25 January 2022 retrieved 1037 references, and searches of included study and related systematic review reference lists identified a further 13 records (totalling 1050 references). The original 1037 references were additionally screened by Cochrane's Screen4Me services, which analyses and removes references from the results set before the manual screening stage using the Classifier and known assessments, and this process excluded 481 references. We then manually screened the 569 remaining records, excluding 435 on the basis of title and abstract, removing one by manual deduplication, and marking 133 for full‐text review. For five of these references, abstracts or full text (or both) were unobtainable through local library and Cochrane interlibrary loan requests, and have been marked as 'awaiting classification' (see Studies awaiting classification). Therefore, 128 studies underwent full‐text screening. Following manual deduplication and grouping of multiple study reports with the primary reference, we excluded a further 61 records, with reasons specified (see Excluded studies). The remaining 36 records were eligible for inclusion.
Included studies
The details of all included studies can be reviewed in the Characteristics of included studies table, with an overview of studies provided in Table 1.
Thirty‐six studies met the inclusion criteria; 35 were double‐blind, randomised, placebo‐controlled trials of a parallel‐group design, including adults classified as having COPD or chronic bronchitis (or both). One trial was of a randomised crossover design; however, outcomes were measured following the initial treatment and follow‐up period, prior to crossover occurring, and, therefore, we included the initial study period in this review (Venge 1996). Two studies were multi‐arm, comparing two intervention groups and placebo (Bonde 1986; De Bernardi 1992). In Bonde 1986, the two intervention groups consisted of different doses of the same immunostimulant agent; in De Bernardi 1992, the two intervention groups consisted of the same immunostimulant agent and dose, but differed by the manufacturing country. Two studies separately examined several participant subsets versus placebo following initial randomisation, respectively creating four and six total comparison groups based on the nature or severity (or both) of the underlying airways disease (Anthoine 1985; Blaive 1982). Two included studies present data in abstract form only; full reports were unable to be obtained (Alvarez‐Sala 2003; Olivieri 2011).
The 36 RCTs recruited 6192 participants. Study duration ranged from three to 14 months, with a mean duration of 6.6 months.
Twelve studies examined the use of immunostimulants in participants with COPD only (Alvarez‐Mon 2005; Alvarez‐Sala 2003; Blaive 1982; Braido 2015; Carlo 1990; Catena 1992; Cazzola 2006; Collet 1997; EUCTR2007‐004702‐27‐DE; Menardo 1985; Olivieri 2011; Xinogalos 1993). The remaining studies included participants with chronic bronchitis or COPD (or both), or chronic bronchitis alone. In studies where included participants were defined as having chronic bronchitis, it was assumed that some or all of the participants had COPD if it was documented in the inclusion criteria or text that they had a degree of airflow limitation or obstruction, or if mean baseline lung function values indicated this.
Two studies indicated that the included participants concurrently had a degree of immunodeficiency (Catena 1992; De Bernardi 1992). Catena 1992 described participants to have 'cell‐mediated immune deficiency' translated to be defined as "Multitest‐Merieux positive with no more than 2 antigens." De Bernardi 1992 indicated presence of 'borderline immunodeficiency' as an inclusion criterion, translated to be diagnosed by Merieux‐Multitest, chemotaxis, phagocytic activity of bronchoalveolar lavage macrophages, or sputum immunoglobulin A (IgA) concentration.
Where studies considered multiple participant groups based on the nature or severity (or both) of airways disease, we only extracted data for the relevant participant subsets. For example, Blaive 1982 presented data separately for participants with asthma, emphysema, and chronic bronchitis; we only extracted data for the emphysema and chronic bronchitis groups. In Anthoine 1985, where participants with chronic bronchitis were grouped into those with (Group II) or without (Group I) airflow limitation, we only extracted data for Group II participants as outcomes for this group were more completely reported by the study authors and deemed of greater relevance to this review.
Inclusion and exclusion criteria
All studies involved adults who fulfilled criteria for chronic bronchitis, COPD, or both. Inclusion and exclusion criteria for each study were variable. Eight studies did not specify any exclusion criteria (Alvarez‐Sala 2003; Blaive 1982; Debbas 1990; Keller 1984; Messerli 1981; Olivieri 2011; Orlandi 1983; Tag 1993), noting that Alvarez‐Sala 2003 and Olivieri 2011 were available in abstract form only. Several did not clarify whether participants with other concomitant respiratory illnesses were excluded. Seven studies specified the inclusion criterion of participants experiencing an acute exacerbation at the time of study enrolment; however, all examined the preventive medium‐ to long‐term (rather than acute treatment) effects of the immunostimulant agent and hence were deemed to be acceptable to include in this review (Cvoriscec 1989; Debbas 1990; Djuric 1989; Keller 1984; Messerli 1981; Soler 2007; Tang 2015).
Lung function
Seventeen studies reported baseline lung function values using FEV1 or FEV1 % predicted. Five of these indicated that the mean FEV1 % predicted of the included participants was less than 50% predicted; eight suggested that on average participants had moderate airflow limitation (mean FEV1 50% or greater and less than 80% predicted); one indicated participants on average had mild airflow limitation (mean FEV1 80% predicted or greater). Three studies reported absolute FEV1 volumes only. Of the remaining studies that did not specify baseline lung function, and excluding Olivieri 2011 (abstract only), 10 included participants with either COPD alone or COPD or chronic bronchitis (or both). The remaining studies were those where included participants had chronic bronchitis, but it was not specified by study authors whether some or all participants may have had a degree of chronic, fixed airflow limitation.
Age
The mean age of participants ranged from 35.2 years (Tag 1993) to 82 years (Orcel 1994). Three studies did not report age (EUCTR2007‐004702‐27‐DE; Olivieri 2011; Rico 1997). Across the remainder of included studies, the mean age of participants was 60.0 years. Of these, 90% of studies included participant groups with a mean age between 45 and 70 years.
Smoking status
Twenty studies reported baseline smoking status. Of the studies that analysed the number of participants who were either active or ex‐smokers, proportions ranged from 27.7% (Menardo 1985) to 95.9% (Bonde 1986). Several studies only reported the number of active smokers, and did not specifically indicate the presence or prevalence of ex‐smokers.
Baseline exacerbation frequency
Twenty‐six studies reported the baseline mean exacerbation frequency, specified a history of frequent respiratory exacerbations as an inclusion criterion, or both.
Thirteen studies reported the baseline mean exacerbation frequency of study participants. Of these, 10 studies indicated participants had a mean exacerbation frequency of two or more in the preceding year (Alvarez‐Mon 2005; Bisetti 1994; Cazzola 2006; Cvoriscec 1989; Fietta 1988; Keller 1984; Menardo 1985; Rochemaure 1988; Soler 2007; Tang 2015). Of the remaining three studies, one reported a mean exacerbation frequency over the preceding two years (Braido 2015), and two reported a mean exacerbation frequency but did not specify the timeframe over which this was measured (Carlo 1990; De Bernardi 1992). However, Carlo 1990 listed an exacerbation frequency of five or greater in the preceding year as an inclusion criterion for enrolment.
A further 13 studies specified a history of respiratory exacerbations as an inclusion criterion but did not report the baseline mean exacerbation frequency. Of these, nine studies stipulated that a history of two or more exacerbations in the preceding year was required for study inclusion (Bonde 1986; Catena 1992; Debbas 1990; EUCTR2007‐004702‐27‐DE; Foschino 1995; Olivieri 2011; Orcel 1994; Venge 1996; Xinogalos 1993). The other four studies listed an inclusion criterion of participants having a history of respiratory exacerbations, but did not specify an exacerbation rate threshold (Anthoine 1985; Ciaccia 1994; Hutas 1994; Li 2004).
Two studies reported variations on the metric of baseline mean exacerbation frequency in the 'baseline characteristics' table. Ciaccia 1994 reported the proportion of participants who had experienced three or fewer or more than three exacerbations over the year preceding the study. Bongiorno 1989 reported the number of acute exacerbation events (rather than participants with events) over the four months preceding the study.
There were no studies that indicated a significant difference in baseline exacerbation frequency between the included intervention and placebo groups.
Immunostimulants and dose
Summary tables regarding variations on the immunostimulant type, dose regimens, and total treatment durations are outlined in Table 1 and Table 2.
| Trade name | Generic name | Active entity |
|---|---|---|
| Adimod | Pidotimod | Synthetic agent |
| Biostim | RU41740 | Bacteria‐derived (glycoproteins and membrane fractions of Klebsiella pneumoniae) |
| Broncho‐Vaxom, Broncho‐Munal, Ommunal, Paxoral, Vaxoral | OM‐85 | Bacterial lysate |
| Diribiotine CK | Not available | Bacterial lysate |
| Hymovis, Monovisc, Orthovisc | Hyaluronan | Synthetic agent |
| Immunoferon, Inmunol | AM3 | Glycophosphopeptical (polysaccharide and protein compounds of Candida utilis) |
| Ismigen | Not available | Polyvalent mechanical bacterial lysate |
| Lantigen B | Not available | Bacterial lysate |
| Ribomunyl, Ribovac, Immucytal | D53 | Bacteria‐derived (proteoglycans of Klebsiella pneumoniae and ribosomal fragments from a range of bacterial pathogens) |
| Thymolin, Leucotrofina | Thymomodulin | Thymic extract |
| Symbioflor | Not available | Bacteria‐derived (components of Enterococcus faecalis) |
In 16 studies, the immunostimulant used was OM‐85, an oral bacterial lysate. Other immunostimulant agents studied included: AM3, an oral polysaccharide/protein compound isolated from Candida utilis (four studies); RU41740 (Biostim), an oral bacterially derived agent consisting of glycoproteins and membrane fractions of Klebsiella pneumoniae (three studies); Ismigen, a sublingual PMBL (two studies); Diribiotine CK, an oral/liquid bacterial lysate (two studies); thymomodulin, an oral thymic extract (two studies); pidotimod, an oral synthetic agent (two studies); D53 (Ribomunyl), a bacterially derived agent consisting of proteoglycans from Klebsiella pneumoniae and various ribosomal fractions, administered in aerosol and subcutaneous form (one study) and orally (one study); Lantigen B, a sublingual bacterial lysate (one study); Symbioflor, an oral/sublingual bacterial lysate (one study); and hyaluronan, a subcutaneous synthetic agent (one study).
Twenty‐six studies included intermittent dosing regimens across the total treatment duration period, and these appeared to be specific to the type of immunostimulant agent. Immunostimulants associated with continuous regimens were AM3 (Alvarez‐Mon 2005; Alvarez‐Sala 2003; Bongiorno 1989; Carlo 1990), thymomodulin (Catena 1992; Rico 1997), pidotimod (Bisetti 1994; Ciaccia 1994), Symbioflor (Habermann 2001), and hyaluronan (Venge 1996).
Across the sixteen studies that analysed OM‐85, there were no variations on the dose of 7 mg/day; however, there were differences in dose pattern and treatment duration, with five studies using a three‐month regimen (Debbas 1990; Li 2004; Messerli 1981; Orcel 1994; Tang 2015), two studies using a four‐month regimen (Collet 1997; Orlandi 1983), and nine studies using a five‐month regimen (Cvoriscec 1989; Djuric 1989; EUCTR2007‐004702‐27‐DE; Hutas 1994; Keller 1984; Olivieri 2011; Soler 2007; Tag 1993; Xinogalos 1993).
In the four studies that included AM3, two used a total daily dose of 3 g/day (Alvarez‐Mon 2005; Alvarez‐Sala 2003), and two used a total daily dose of 1.5 g/day (Bongiorno 1989; Carlo 1990).
For the three studies that included RU41740 (Biostim), two used a dose of 2 mg/day for an initial interval followed by 1 mg/day for subsequent intervals (Anthoine 1985; Fietta 1988). One was multi‐arm and used 2 mg/day for the first intervention group and 8 mg/day for the second intervention group (Bonde 1986).
For the two studies that included thymomodulin, one used a total daily dose of 120 mg/day (Catena 1992), and the other used a total daily dose of 240 mg/day (Rico 1997).
For the two studies that included Ismigen, both used a total daily dose of 50 mg/day; however, one study used an intermittent dosing regimen over nine months' total duration (Braido 2015), and the other over three months' total duration (Cazzola 2006).
In the two studies that included Ribomunyl, the individual medication doses were not specified and there was variation between both the dose regimens and the method of administration, with Blaive 1982 using a combination of aerosol and subcutaneous routes and Foschino 1995 using an oral form.
Study size and duration
Study size ranged from 19 participants (Orlandi 1983) to 494 participants (Ciaccia 1994). Total study duration ranged from three months (Alvarez‐Sala 2003; Carlo 1990; Catena 1992) to 14 months (Habermann 2001) with a mean duration of 6.6 months. Total treatment duration, which included the time intervals between periods of treatment where intermittent dosing regimens were used, ranged from two months (Bisetti 1994; Ciaccia 1994) to nine months (Braido 2015) with a mean duration of four months.
Countries
Ten studies were conducted in Italy; five in France; three in Switzerland; two in Spain; two in countries of the former Yugoslavia; two in China; and one each in Canada, Mexico, Egypt, Greece, Hungary and Germany. Six studies were conducted across two or more European/Scandinavian countries.
Funding
Nine studies reported pharmaceutical sponsorship (Alvarez‐Mon 2005; Blaive 1982; Braido 2015; Collet 1997; Cvoriscec 1989; EUCTR2007‐004702‐27‐DE; Hutas 1994; Tang 2015; Venge 1996). Five studies did not specifically state pharmaceutical sponsorship; however, authorship details or the listed study address (or both) suggest pharmaceutical company associations (Bonde 1986; Menardo 1985; Messerli 1981; Rico 1997; Soler 2007). Fietta 1988 reported partial sponsorship from public grant funding. The remaining 21 studies did not specify any sponsorship or funding sources.
Excluded studies
We excluded 92 studies after full‐text screening. See Characteristics of excluded studies table for reasons for exclusion.
Risk of bias in included studies
Details of our risk of bias judgements are presented in the risk of bias section of the Characteristics of included studies table, with an overview in Figure 2.
Allocation
All included studies were reported to be randomised. However, for most studies the potential for allocation bias was unclear for both random sequence generation and allocation concealment, in that the authors did not state in sufficient detail the method of randomisation, where this took place, and how it was concealed. One study was at low risk of bias for both sequence generation and allocation concealment (Braido 2015). One study was at low risk for allocation concealment, but unclear for sequence generation (Orcel 1994). One study was at high risk for both domains (Orlandi 1983).
Of the studies that were unclear regarding potential for allocation bias, 21 did not provide any information regarding the randomisation process or how concealment was achieved. Five studies used random‐draw, randomly planned sequences, or randomisation codes, but gave no further details (Anthoine 1985; Blaive 1982; Bongiorno 1989; Messerli 1981; Rochemaure 1988). Six studies used block permutation or stratified randomisation methods (or both), but it was unclear how the sequences were generated (Alvarez‐Mon 2005; Bonde 1986; Ciaccia 1994; Collet 1997; Orcel 1994; Soler 2007). One study indicated randomisation by a computer‐generated sequence without further detail (Ciaccia 1994). One study did not specify methods within the published report, but following further contact with the study author it was indicated computer‐generation may have been used (Hutas 1994).
Two studies involved unbalanced randomisation in the intervention versus placebo group (Catena 1992: 2:1; Tag 1993: 3:2).
Most studies reported the baseline characteristics of treatment groups, which appeared well‐matched. The authors of several studies only reported characteristics for the overall study population (Bonde 1986; Fietta 1988; Orlandi 1983), or sparse numerical data regarding baseline characteristics (Djuric 1989; Messerli 1981), but commented that the comparison groups were homogeneous for a range of metrics. One study provided data for its overall study population but did not specify whether the groups were well‐matched (Venge 1996). Two studies did not report on population baseline characteristics (EUCTR2007‐004702‐27‐DE; Olivieri 2011), although Olivieri 2011 was presented as an abstract only.
Blinding
All included studies were reported to be double‐blind. However, most did not provide sufficient information regarding the methods used to achieve blinding or the groups that were blinded; therefore, the potential of bias for these domains was unclear for 34 studies. One study was at low risk of bias for blinding of participants and personnel, but whether outcome assessors were blinded was not specified (Braido 2015). One study was at low risk of blinding of outcome assessors, but unclear for blinding of participants and personnel (Collet 1997).
Authors of eight studies indicated that the intervention and placebo formulations (tablets, capsules, or drops) were identical in appearance (Carlo 1990; Collet 1997; De Bernardi 1992; Djuric 1989; Keller 1984; Messerli 1981; Orcel 1994; Xinogalos 1993). Whilst noted as an important feature of ensuring adequate participant and personnel blinding, if there was no elaboration on methodology to ensure blinding or the groups that were blinded (or both) were not specified, then this information alone was deemed insufficient to create confidence that there was adequate protection against performance and detection bias.
Incomplete outcome data
Study dropout rate varied from a reported 0% (Carlo 1990) to an estimated, but not explicitly stated, 72% (Bisetti 1994). When the proportion of dropouts was more than 20% of the total study population, we considered a 'high risk' rating; however, it was taken into account whether the dropouts were balanced in number and had occurred for similar reasons between study arms. The proportion of dropouts relative to study duration, and whether an ITT versus completer analysis had been performed, were also considerations.
Five studies were at high risk of attrition bias (Alvarez‐Sala 2003; Bisetti 1994; EUCTR2007‐004702‐27‐DE; Menardo 1985; Xinogalos 1993), noting that Alvarez‐Sala 2003 was presented in abstract form only. In EUCTR2007‐004702‐27‐DE, study authors reported their concerns about significant bias in this domain due to large amounts of missing data and flaws in analysis contributed to by invalid and inaccurate assessments of outcome events. As a result, the study authors and an independent group of experts deemed the study to be flawed and efficacy conclusions unable to be made based on the available study results. It may be relevant that this was a pharmaceutical company‐sponsored trial where the results for the primary outcome were unfavourable for the intervention group compared to placebo. Further information about the reported processes that led to this conclusion can be viewed in the Characteristics of included studies table.
Seven studies were at low risk of attrition bias, as dropout rates were low or had been adequately described such that they were unlikely to have contributed to the quality and interpretation of study outcomes (Anthoine 1985; Braido 2015; Carlo 1990; Cazzola 2006; Collet 1997; Orlandi 1983; Tang 2015).
The remaining studies were at unclear risk. An 'unclear' rating was given if dropout rates, reasons, distribution across the comparison groups, or a combination of these were not well‐described or if dropouts were unbalanced between groups. If dropout rates were high and per‐protocol analysis was undertaken, even if well‐balanced between groups and reasons provided, these studies were also deemed at unclear risk.
Six studies performed an ITT analysis (Anthoine 1985; Braido 2015; Cazzola 2006; Collet 1997; Debbas 1990; Tang 2015), and one study analysed the full participant population due to a reported 100% completion rate (Carlo 1990). Several studies did not describe the presence or absence of dropouts, but the number of participants analysed at study completion appeared to match those randomised at study commencement; therefore, it was not clear whether an ITT analysis was incorporated with dropouts or exclusions having occurred, or whether all participants had completed the study (Cvoriscec 1989; De Bernardi 1992; Djuric 1989; Habermann 2001; Li 2004; Rico 1997; Venge 1996). The remaining 22 studies used a per‐protocol analysis. Of these, one study indicated the allocation of an ITT sample based on its study flow diagram; however, they used a per‐protocol analysis for the primary study outcomes (Soler 2007).
Selective reporting
One study was at high risk for bias in this domain, due to methodological and outcome assessment flaws that were identified and reported by the study authors and an independent review panel (EUCTR2007‐004702‐27‐DE; see Characteristics of included studies table). Excluding the two studies presented as abstract only (Alvarez‐Sala 2003; Olivieri 2011), nine studies were at unclear risk, because not all intended outcomes were reported, there was possible skew towards reporting of positive outcomes, or data had been presented in such a way that may have positively influenced the significance of the findings (Anthoine 1985; Bonde 1986; Ciaccia 1994; Debbas 1990; Foschino 1995; Menardo 1985; Orlandi 1983; Rochemaure 1988; Xinogalos 1993). The remaining 24 studies appear to have reported sufficiently on all intended outcomes and were, therefore, judged at low risk of bias in this domain.
Other potential sources of bias
One study indicated that baseline data for outcomes measuring the efficacy of the intervention was obtained through retrospective questionnaire, which may have introduced hindsight bias (Rico 1997). The impact of this on study outcomes was unclear, as baseline data did not appear to be relevant to primary or secondary outcome analysis.
Effects of interventions
See: summary of findings Table 1 for an overview of the main results together with a summary of our confidence in the evidence per outcome.
Primary outcome: number of participants with no exacerbations during the study period
Sixteen studies reported this outcome directly or indirectly (by reporting the number of participants with an exacerbation as a number of proportion of the total group population, from which the number of participants with no exacerbations could be inferred). Fifteen were included in meta‐analysis; one study was not included, as outcomes were analysed over two discrete eight‐week time intervals and this duration was shorter than had been specified in our protocol criteria (Bisetti 1994).
Immunostimulant medication increased the overall odds of not experiencing an exacerbation over the study period compared to placebo (OR 1.48, 95% CI 1.15 to 1.90; I2 = 53%; 15 studies, 2961 participants; Analysis 1.1; Figure 3; moderate‐certainty evidence). For the mean comparator risk of 0.52, this corresponded to an NNTB of 11 (95% CI 7 to 29). For the extremes of low (0.05) comparator risks for this outcome, the NNTB was 46 (95% CI 25 to 143) and for high (0.68) comparator risks was 13 (95% CI 9 to 34).
The heterogeneity across the 15 studies included in this analysis was moderate to high (I2 = 53%), and was explored using preplanned subgroup analyses. Subgrouping for immunostimulant type was not associated with a consistent reduction of heterogeneity within subgroups, and the test for subgroup differences indicated that there was no evidence of a subgroup treatment effect (Chi2 = 2.82, degrees of freedom (df) = 3 (P = 0.42), I2 = 0%; Analysis 1.2). Preplanned subgroup analyses were also undertaken to assess for any variations in treatment effect related to baseline lung function and exacerbation frequency. Results of the subgroup and sensitivity analyses performed are elaborated on in 'Subgroup and sensitivity analyses' below.
The funnel plot indicated the absence of small, neutral, or negative studies, and the likely presence of publication bias which may have skewed the pooled effect estimate towards a more positive result (Figure 4). However, exclusion of the five small, positive studies by sensitivity analysis did not significantly lessen the effect estimate.
Two included studies measured this outcome in discrete time intervals (Habermann 2001; Menardo 1985). Unlike Bisetti 1994 however, the data sets from the initial time period were incorporated into meta‐analysis as they had been measured over longer intervals of three months (Menardo 1985) and six months (Habermann 2001). The later data sets were not included in meta‐analysis. The data set for the first, rather than the last, measurement period from baseline was preferentially incorporated, as it was considered that the study participants were more likely to be truly randomised at study commencement and there would be less risk of a carry‐over effect from earlier treatment exposure, which may potentially confound outcome data.
Bisetti 1994 (pidotimod) measured outcomes over four months in discrete two‐month intervals. Authors reported that fewer participants experienced exacerbations in the intervention group compared to placebo for both the zero‐ to two‐month interval (9/88 participants with intervention versus 57/75 participants with placebo) and the three‐ to four‐month interval (0/25 participants with intervention versus 13/26 participants with placebo) (reported P < 0.001 for both time intervals). However, concerns around methodology including significantly high and unexplained dropout rates led to a 'high risk' judgement for attrition bias, and the calculated ORs and CIs for the inverse of these results (i.e. the number of participants with no exacerbations) also suggested data imprecision for both the zero‐ to two‐month (OR 27.80, 95% CI 11.65 to 66.32; 163 participants) and three‐ to four‐month (OR 51.0, 95% CI 2.81 to 925.71; 51 participants) study intervals.
Menardo 1985 (Diribiotine CK), measured number of participants with no exacerbations over six months in two discrete three‐month intervals. The zero‐ to three‐month data set was included in meta‐analyses. Due to concerns regarding attrition bias, this study was at 'high risk' for this domain; however, exclusion by sensitivity analysis did not lead to augmentation of the pooled effect estimate. The four‐ to six‐month data set was not included, but for this later time period the study authors reported a difference in the number of participants without an exacerbation in the intervention group compared to placebo (5/10 participants with intervention versus 2/10 participants with placebo; OR 3.5, 95% CI 0.47 to 25.90; P = 0.22). Data imprecision lowers the certainty in this effect estimate.
Habermann 2001 (Symbioflor) measured number of participants with no exacerbations over 14 months in discrete six‐ and eight‐month intervals. The zero‐ to six‐month data set was included in meta‐analysis. Results for the seven‐ to 14‐month, post‐treatment follow‐up period indicated that the number of participants with an exacerbation during this time was less in the intervention group compared to placebo (P = 0.013, as reported by the study authors). The associated point estimate of effect suggested an increased odds of experiencing no exacerbations with intervention compared to placebo (OR 1.83, 95% CI 0.9 to 3.7; 136 participants); however, the estimate is imprecise.
Sixteen other studies that did not report the number of participants with or without an exacerbation, instead reported on alternative exacerbation metrics such as the number of exacerbation events or the mean exacerbation rate for the study period. The studies varied in the way they presented count data in terms of method of presentation, reporting of variance, and rate time frames. Therefore, this information was not extracted or pooled for the purposes of this systematic review.
Primary outcome: mortality (all‐cause)
Seven studies (2003 participants) reported mortality data (Bonde 1986; Braido 2015; Cazzola 2006; Collet 1997; EUCTR2007‐004702‐27‐DE; Orcel 1994; Soler 2007). Five were combined in meta‐analysis. We did not include Bonde 1986 as they reported a 1.2% overall mortality rate (172 participants), but it was not specified which comparison groups the deaths were associated with, and Soler 2007 reported zero‐event rates in both arms (273 participants).
Immunostimulants probably result in little to no difference in all‐cause mortality compared to placebo, measured over a mean follow‐up period of eight months (OR 0.64, 95% CI 0.37 to 1.10; I2 = 0%; 5 studies, 1558 participants; Analysis 1.8; Figure 5; moderate‐certainty evidence); however, CIs were wide and included the potential for both a clinically important difference and no difference.
The studies that were meta‐analysed only included two immunostimulant agents; OM‐85 (three studies) and Ismigen (two studies).
Orcel 1994 contributed most of the weight in the meta‐analysis for all‐cause mortality and recruited elderly participants from residential care facilities (mean age 82 years) with a diagnosis of chronic bronchitis. Reported baseline characteristics data suggested high rates of comorbidity in both groups. Therefore, the results from this study may be less applicable to the general chronic bronchitis or COPD (or both) population; however, there was little impact on the pooled effect estimate when this study was excluded in sensitivity analysis.
Secondary outcome: mortality (respiratory‐related)
Two larger studies that had reported on all‐cause mortality also reported respiratory‐related mortality (Collet 1997; Orcel 1994). Both analysed the immunostimulant OM‐85 over six months. Results from pooled analysis indicated there may be little to no difference in respiratory‐related mortality (OR 0.40, 95% CI 0.15 to 1.07; I2 = 0%; 2 studies, 735 participants; Analysis 1.14; low‐certainty evidence). However, data availability, imprecision, and indirectness lower the certainty in this effect estimate. The concerns regarding the applicability and generalisability of the data presented in Orcel 1994, with its inclusion of an elderly, comorbid study population, carried greater weight for this outcome considering this meta‐analysis included only two studies.
Secondary outcome: quality of life
Although several studies reported subjective participant or physician (or both) estimations of well‐being, only six explored the impact of immunostimulants on health‐related quality of life (HRQoL) compared to placebo using validated assessment tools. The tools used included the SGRQ (Jones 1992), the 36‐item Short Form Health Survey (SF‐36), the 12‐item Short Form Health Survey (SF‐12), and the Chronic Cough Impact Questionnaire (CCIQ) (Baiardini 2005).
In three studies, investigators assessed the impact of the intervention on quality of life using the SGRQ (Alvarez‐Mon 2005; Alvarez‐Sala 2003; EUCTR2007‐004702‐27‐DE). Collet 1997 used the SF‐36. Braido 2015 used both the SF‐12 and CCIQ. Catena 1992 used an unnamed, 9‐point 'Index of improvement for quality of life' scale that referenced an earlier study in which the scale appeared to have been validated (Grossi 1989); the original publication of this article could not be obtained.
The SGRQ is made up of three subscales – symptoms, activities, and impacts – to yield a total score ranging from 0 to 100 (Jones 1992). A lower score indicates a better quality of life. Total SGRQ scores from two studies, both of which assessed the immunostimulant AM3, were combined in meta‐analysis (Alvarez‐Mon 2005; Alvarez‐Sala 2003); scores from EUCTR2007‐004702‐27‐DE could not be incorporated as there were no numerical data presented, with authors reporting that there were no differences between study arms.
There was a reduction in total SGRQ scores for participants receiving immunostimulant compared to placebo (MD −4.59, 95% CI −7.59 to −1.59; I2 = 0%; 2 studies, 617 participants; Analysis 1.15; very low‐certainty evidence). This effect met the minimum clinically important difference (MCID) of a reduction of 4 units on the SGRQ scale (Jones 2005). However, the upper limit of the CI for this effect did not clear the MCID and the possibility of a non‐meaningful difference could not be excluded. Several other factors contributed to a lowering of the certainty of this effect estimate; Alvarez‐Sala 2003 was presented as an abstract only, had a short duration of follow‐up (three months), and, due to inconsistencies in the reported number of participants for each arm, was at high risk for attrition bias. Additionally, there were no standard deviations for the total scores in each study arm reported and, for the purposes of meta‐analysis these were instead calculated from the reported P value for the MD between comparison groups.
Alvarez‐Mon 2005 reported both mean total SGRQ scores and mean change‐from‐baseline scores over the six‐month study period for 253 participants. There was a difference in the total SGRQ scores (MD −4.6) and total 'activity' subcomponent scores (MD −7.1) described between the comparison groups at six months, favouring the intervention (P < 0.05 for both outcomes, as reported by study authors). The differences between study arms with respect to the total 'symptom' (MD −3.4) and 'impact' (MD −4.1) subcomponent scores, although favouring intervention, were reportedly non‐significant. In assessing change in mean scores from baseline, there were no differences for total SGRQ (MD −3.3; reported P = 0.076), 'activity' (MD −1.7), and 'impact' (MD −3.1) subcomponent scores, comparing intervention to placebo. However, there was a meaningful difference favouring intervention between study arms in the change of mean 'symptom' subcomponent scores from baseline (MD −5.7; reported P < 0.05).
For the three studies that analysed the effect of an immunostimulant agent on quality of life using alternative assessment tools, there were no numerical data were presented by trialists to enable qualitative or quantitative analysis in this review (Braido 2015; Catena 1992; Collet 1997). In Braido 2015 (Ismigen; 12 months' follow‐up; 288 participants) and Collet 1997 (OM‐85; six months' follow‐up; 381 participants), the study authors reported no differences in the quality‐of‐life scale scores between intervention and placebo groups, with data not shown. In Catena 1992 (thymomodulin; three months' follow‐up; 236 participants) authors provided a graphical representation of nine quality‐of‐life domains and presented data per domain for both intervention and placebo groups as a ratio of percentage of improved participants to percentage of participants who were 'worse' at the completion of the study per domain. They reported improvement in all nine domains for the intervention group and in four domains for the placebo group (reported P < 0.05). There was no comparison between the intervention and placebo groups for each domain or overall reported.
Secondary outcome: number of participants requiring antibiotics
Seven studies reported number of participants requiring antibiotics; five were combined in meta‐analysis (Anthoine 1985; Debbas 1990; De Bernardi 1992; Habermann 2001; Menardo 1985). Two studies could not be included as outcomes were analysed in discrete monthly time intervals, and this duration was shorter than our specified protocol criteria (Cvoriscec 1989; Xinogalos 1993). The studies combined in meta‐analysis included the immunostimulant agents RU41740 (Biostim) (one study), OM‐85 (one study), Lantigen B (one study), Symbioflor (one study), and Diribiotine CK (one study). The two studies not meta‐analysed both examined the effects of the immunostimulant OM‐85 BV.
Immunostimulants likely reduced the overall odds of receiving antibiotic treatment for an exacerbation over the study period compared to placebo (OR 0.34, 95% CI 0.18 to 0.63; I2 = 38%; 5 studies, 542 participants; Analysis 1.16; Figure 6; moderate‐certainty evidence). Although the upper limit for the risk ratio CI crossed the 25% relative risk threshold for appreciable benefit (RR 0.64, 95% CI 0.5 to 0.82), the optimal information size criteria was met and the CI did not include the null effect.
The heterogeneity of these five studies was low to moderate, and likely due to the clinical and methodological diversity between studies, noting that each included study examined a different immunostimulant agent. However, the direction of effect was consistent across studies. One studies included in the analysis was at high risk for attrition bias (Menardo 1985). Another study included participants with COPD and concurrent 'borderline immune deficiency' and there were concerns about the applicability of these data as a result (De Bernardi 1992). Another examined an elderly population (mean age 81.8 years), which may affect applicability and generalisability of these data to a general chronic bronchitis/COPD population (Debbas 1990). However, individual exclusion of these studies by sensitivity analyses had little impact on the pooled effect estimate.
One study included in the meta‐analysis measured number of participants requiring antibiotics over six months in two discrete three‐month intervals (Menardo 1985). We incorporated only the initial zero‐ to three‐month data set. The study authors also presented data on the number of participants not requiring antibiotics over the four‐ to six‐month period. The inverse of the reported results for this interval (i.e. the number of participants requiring antibiotics) corresponded to an OR of 0.64 (95% CI 0.11 to 3.92; P = 0.65; 20 participants); however, this estimate is imprecise.
Two studies analysed the effect of immunostimulant on the number of participants requiring antibiotics compared to placebo over six months in discrete monthly intervals (Cvoriscec 1989; Xinogalos 1993). Cvoriscec 1989 (104 participants) presented data in graphical form, and authors only reported on the reduction in the proportion of participants requiring antibiotics compared to study entry in the intervention group (reported P < 0.05). There were no statistical comparisons of the intervention and placebo groups presented for this outcome at any of the study time points. Xinogalos 1993 (62 participants) reported that for the first three months of the trial there were no differences in the number of participants requiring antibiotics between the study arms, with numerical data not shown. They reported data for the four‐ and six‐month time intervals, with differences noted in the proportion of participants requiring antibiotics in the immunostimulant group compared to placebo (four months: P < 0.001; six months: P < 0.002; reported by the authors). For both studies, it was difficult to draw conclusions about the impact of immunostimulants on the requirement for antibiotic use, as the short time intervals over which this outcome was measured were considered likely to overestimate treatment effect. Additionally, the high risk of attrition bias in Xinogalos 1993 impacted interpretation of data quality and certainty regarding the effect estimates.
Several studies that did not report this outcome presented data regarding antibiotic use by alternative metrics; for example, as duration of antibiotic therapy (eight studies) or the incidence of antibiotic use (three studies).
Secondary outcome: exacerbation duration
Twenty‐one studies analysed exacerbation duration. Seventeen studies were initially included in meta‐analysis; three studies reported results over discrete intervals that did not meet our protocol criteria for minimum duration of outcome measurement (Bongiorno 1989; Ciaccia 1994; Xinogalos 1993). One study reported that there were no differences in this outcome between comparison groups, with no numerical data shown (EUCTR2007‐004702‐27‐DE).
Of the studies included in the meta‐analysis, one measured outcomes over 12 months in discrete three‐month intervals (Rico 1997); we incorporated only the first zero‐ to three‐month period from baseline.
The meta‐analysis performed for these 17 studies (1693 participants) indicated a pooled effect estimate that favoured intervention, measured over a mean duration of 6.4 months; however, this result was associated with an unacceptably high level of heterogeneity (I2 = 92%; Analysis 1.17). Therefore, the mean effect was difficult to interpret and draw conclusions from, and the likelihood of data skew was also considered to make parametric analysis less robust. Consensus decision was made to disable the forest plot totals, although the graph has still been included in this review for transparency and reference purposes.
The degree of heterogeneity seen was interpreted to be due to methodological and clinical diversity related to study interventions, design, population (noting variability in smoking prevalence and mean participant age between included studies), potentially variable outcome definitions and measures of effect, and the use of a continuous variable. The variability in absolute values occurring between studies may indicate that results could have been presented by authors as either the mean duration of an exacerbation episode, or as the total or cumulative number of exacerbation days averaged over the number of participants for the study duration. For several studies, this differentiation was unclear. In five studies measures of variance had not been reported by the trialists; these were instead calculated and assumed from the reported MDs and P values (Cazzola 2006; Foschino 1995; Olivieri 2011; Orlandi 1983; Tag 1993). A sensitivity analysis, excluding these studies and separately those that that were at high risk of bias for any domain, did not significantly modify the pooled effect estimate.
In lieu of formally meta‐analysing this outcome, we performed a post hoc synthesis of data by vote‐counting based on direction of effect, whereby the effect estimate for each study was categorised as a binary metric (beneficial versus harmful) and a binomial probability test applied to determine the probability of observing the result if the intervention was truly ineffective (Figure 7). Of the 18 available studies, one provided no information about direction of effect and was excluded (EUCTR2007‐004702‐27‐DE). By this method, there was evidence that immunostimulant agents had an effect on reducing exacerbation duration, with 16/17 included studies favouring the intervention (94%, 95% CI 73% to 99%; P = 0.0002). Sensitivity analyses that excluded studies at high risk of bias in any domain or for which missing variance data had been imputed did not impact significantly on these results.
Effect direction plot and sign test for the outcome of exacerbation duration.
Rico 1997 (thymomodulin; 88 participants) measured the impact of exacerbation duration over 12 months in discrete three‐month intervals. We included the zero‐ to three‐month data set in the above analyses, but not the later time periods. Authors reported reductions in exacerbation duration in the intervention group compared to placebo during the four‐ to six‐month (MD −2.09; reported P = 0.04), seven‐ to nine‐month (MD −1.26; reported P = 0.15) and 10‐ to 12‐month (MD −0.49; reported P = 0.44) intervals.
Bongiorno 1989 (AM3; 40 participants) analysed exacerbation duration over four months in discrete two‐month intervals. They reported a longer exacerbation duration in the intervention group versus placebo in the zero‐ to two‐month period (MD 1.65), and conversely a reduction in duration favouring immunostimulant over the three‐ to four‐month period (MD −2.45; reported P < 0.005).
Ciaccia 1994 (pidotimod) measured exacerbation duration over five months in two‐ and three‐month discrete intervals. Authors presented data for all participants (492 participants at two months; 481 at five months), and also for those with a history of three or fewer exacerbations in the preceding year (242 participants two months; 240 at five months). For all participants, there was a reduced exacerbation duration in the intervention group compared to placebo for both time periods, but authors reported this difference was only significant in the second, three‐month period (MD −2.2; reported P < 0.01). For those who were less‐frequent exacerbators, there was a difference favouring intervention across both time periods (MD −1.6 for zero to two months; and MD −2.2 for three to five months; reported P < 0.05 for both intervals). There were no data for those who were more‐frequent exacerbators (greater than three exacerbations in the preceding year).
Xinogalos 1993 (OM‐85; 62 participants) analysed the impact of immunostimulant on exacerbation duration over six months, in one‐month intervals. They reported numerical data for all time periods. In the first and third months, exacerbation duration was less in the intervention groups compared to placebo, but this difference was reported to be non‐significant. In the second month, the mean exacerbation duration was higher in the intervention group (MD +0.79). For the remainder of the study, authors reported a reduction in the mean exacerbation duration in the intervention group compared to placebo when measured over the fourth (MD −1.14; reported P < 0.001), fifth (MD −0.52; reported P < 0.05), and sixth months (MD −0.16; P < 0.002).
Secondary outcome: hospitalisation duration (respiratory‐related)
Seven studies analysed the effect of immunostimulant on hospitalisations (either all‐cause or respiratory‐related) compared to placebo (Braido 2015; Cazzola 2006; Collet 1997; EUCTR2007‐004702‐27‐DE; Keller 1984; Orlandi 1983; Rico 1997). Three of these reported mean respiratory‐related hospitalisation duration; however, data were unable to be meaningfully combined in meta‐analysis (Cazzola 2006; Collet 1997; Orlandi 1983). A forest plot was initially created combining the data on 559 participants from Cazzola 2006 and Collet 1997 (Analysis 1.18), but heterogeneity was high (I2 = 83%) and the pooled effect estimate difficult to interpret, therefore, totals were disabled. Orlandi 1983 could not be included as the trialists did not report measures of variance, and these were not estimable using alternative statistical methods from the data provided.
Collet 1997 (OM‐85; 381 participants) reported that the mean duration of respiratory‐related hospitalisation was shorter in the immunostimulant group (6.4 days) than in the placebo group (11.3 days) over six‐month follow‐up, although this difference was not significant (MD −4.9, 95% CI −7.47 to −2.33; reported P = 0.058). Cazzola 2006 (Ismigen; 178 participants) reported a reduced mean respiratory‐related hospitalisation duration in the immunostimulant group compared to placebo over 12‐month follow‐up (MD −1.6, 95% CI −2.5 to −0.7; reported P < 0.05). Orlandi 1983 (OM‐85; 19 participants) reported a mean hospitalisation duration of 0.8 days in the intervention group over four months versus 1 day in the placebo group, but did not report measures of variance or statistical significance.
Seven studies measured the impact of immunostimulant therapy on hospitalisations using a range of other metrics. These included the number of respiratory‐related hospitalisation events (three studies), number of all‐cause hospitalisation events (two studies), number of participants requiring hospitalisation for a respiratory cause (one study), number of participants requiring hospitalisation for any cause (one study), total respiratory‐related hospitalisation days (three studies), total all‐cause hospitalisation days (two studies), and mean number of hospitalisation events per participant (one study). The three studies (847 participants) that reviewed total respiratory‐related hospitalisation days were larger trials conducted over a mean follow‐up of 10 months and examined the immunostimulants OM‐85 and Ismigen. All reported decreases in total hospitalisation time in the intervention groups compared to placebo (percentage difference: Braido 2015: Ismigen, −60%; Cazzola 2006: Ismigen, −53%; Collet 1997: OM‐85, −55%).
Secondary outcome: adverse events/side effects
Twenty‐seven studies reported the proportion of participants who experienced an adverse event over the study period. Seven, involving a total of 955 participants, reported no events in either study arm (Carlo 1990; Cazzola 2006; Debbas 1990; De Bernardi 1992; Fietta 1988; Orcel 1994; Venge 1996). The remaining 20 studies were included in meta‐analysis.
There was no evidence of a difference in the odds of experiencing an adverse event when receiving immunostimulant compared to placebo (OR 1.01, 95% CI 0.84 to 1.21; I2 = 0%; 20 studies, 3780 participants; Analysis 1.19; high‐certainty evidence). The CIs included the null effect, but optimal information size criteria were met and the risk ratio upper limit did not cross the relative risk threshold for appreciable harm (RR 1.02, 95% CI 0.92 to 1.13).
Of the remaining studies, Collet 1997 reported the number of adverse event occurrences between comparison groups (138 events in 191 participants with intervention versus 151 events in 190 participants with placebo). Blaive 1982 reported treatment intolerance in 2% of 184 participants overall but did not specify which groups the affected participants belonged to. Bonde 1986 reported 51 cases of 'mild' adverse reaction events in 172 participants overall and did not specify in which groups these events occurred, although authors stated that there were no significant differences between the study arms. Six studies did not report adverse events (Alvarez‐Sala 2003; Bongiorno 1989; Djuric 1989; Hutas 1994; Keller 1984; Xinogalos 1993).
The nature of reported adverse events varied between studies, but in both intervention and placebo groups more frequently included gastrointestinal upset (nausea, dyspepsia, diarrhoea, constipation, abdominal pain), skin itch or rash, upper or lower respiratory tract symptoms, and fever. Eight studies reported study withdrawals that were related to the development of adverse events/side effects; however, these were balanced across the comparison groups (Alvarez‐Mon 2005; Bisetti 1994; Braido 2015; Catena 1992; Ciaccia 1994; Messerli 1981; Rochemaure 1988; Soler 2007). In the remaining studies, there were either no dropouts that occurred due to adverse events or the trialists did not report this.
Subgroup and sensitivity analyses
Subgroup analyses
Planned subgroup analyses were undertaken for the primary outcomes only; that is, participants with no exacerbations and mortality (all‐cause), to investigate for any potential heterogeneity encountered and differences in treatment effect. Subgrouping was performed by immunostimulant type, severity of COPD based on lung function testing (FEV1 50% or greater versus less than 50%), mean baseline exacerbation frequency (two or more in the preceding year versus fewer than two in the prespecified year or unspecified), treatment duration (three months or less versus greater than three months), and study duration (three to less than six months versus six months to less than 12 months versus 12 months or greater).
For the number of participants with no exacerbations, the heterogeneity of the 15 studies included in meta‐analysis was moderate (I2 = 53%). Heterogeneity did not consistently appear to be lower within any of the subgroups than across all studies (Analysis 1.2; Analysis 1.3; Analysis 1.4; Analysis 1.5; Analysis 1.6); where subgroups had no or low levels of heterogeneity suggested by the I2 statistic, the associated P value for the Chi2 test was > 0.1.
To further explore possible sources of heterogeneity associated with this outcome, we performed a post hoc subgroup analysis differentiating studies by the decade of publication (Analysis 1.7). Moderate‐to‐high levels of heterogeneity remained within the subgroups, particularly across the studies published between 1990 and 1999 (I2 = 70%) and 2000 to 2009 (I2 = 66%).
In terms of the potential impact of subgrouping on estimates of treatment effect, there were no significant subgroup effects found and the test for subgroup differences across all subgroup analyses was negative, for both primary outcomes. However, for the outcome of participants with no exacerbations, as there was moderate to substantial unexplained heterogeneity demonstrated between trials within most subgroups, the validity of the treatment effect estimate for each subgroup and the true presence or absence of subgroup differences was uncertain. Additionally, in the subgroup analyses of immunostimulant type (Analysis 1.2), lung function (Analysis 1.3), study duration (Analysis 1.6), and decade of publication (Analysis 1.7), several subgroups contained a limited number of studies; therefore, uneven covariate distribution also limits certainty of the impact of subgrouping effect. For all‐cause mortality, this is also a concern, with low minimum (two) and maximum (three) subgroup study numbers.
When the number of participants with no exacerbations was subgrouped by decade of publication, visualisation of the forest plot implied a trend towards the production of more conservative results in studies that were more recently published. Studies published between 2010 and 2019 (OR 1.08, 95% CI 0.79 to 1.48; I2 = 0%; 2 studies, 672 participants) or between 2000 and 2009 (OR 1.27, 95% CI 0.98 to 1.66; I2 = 66%; 4 studies, 1002 participants) appeared to have a lesser effect size than those published between 1990 and 1999 (OR 1.45, 95% CI 1.12 to 1.88; I2 = 70%; 4 studies, 965 participants) and before 1990 (OR 2.15, 95% CI 1.26 to 3.66; participants 322; studies 5; I2 = 25%; 5 studies, 322 participants). However, inferences about effect sizes were limited by the degree of heterogeneity within the subgroups and uneven covariate distribution.
Sensitivity analyses
In accordance with our protocol, sensitivity analyses were performed excluding studies at high risk of bias in any of the six domains, for any outcome. Where sensitivity analyses were undertaken, this is specified above under the relevant outcome headings. This type of sensitivity analysis did not lead to appreciable modification of the pooled effect estimate for any of the review outcomes.
Post hoc sensitivity analyses were also performed excluding trials that were more likely to lessen the overall data quality, to determine the impact of their absence on the effect estimate. These analyses are also specified above under the relevant outcome headings. Where these were conducted, this was little to no impact on the effect estimates observed.
Last, planned sensitivity analyses were undertaken for outcomes where there had been assumptions concerning missing data. This was largely relevant for the HRQoL and exacerbation duration outcomes, where missing variance measures for individual studies had been calculated from reported MDs and P values. As the meta‐analysis for HRQoL only contained two studies, conclusions were unable to be drawn from a sensitivity analysis that excluded the study where data assumptions had been made (Alvarez‐Sala 2003). The meta‐analysis for exacerbation duration was not undertaken for aforementioned reasons; instead, this outcome was analysed dichotomously using a direction‐of‐effect table (Figure 7). Sensitivity analyses were performed excluding the studies at high risk of bias in any domain or those where data had been assumed, both of which had little impact on the overall proportion of 'positive' studies or associated binomial probability.
Discussion
Summary of main results
The review included 36 studies involving 6192 participants. Studies were published between 1981 and 2015. Study duration ranged from three to 14 months. Immunostimulants investigated were OM‐85, AM3, RU41740 (Biostim), Ismigen, Diribiotine CK, thymomodulin, pidotimod, D53 (Ribomunyl), Lantigen B, Symbioflor, and hyaluronan, with administration including oral, sublingual, and subcutaneous routes. Twenty‐six studies included intermittent, rather than continuous, dosing regimens. The risks of bias of the included studies were mostly low or unclear.
The mean age of study participants ranged between 35.2 and 82 years. Twelve studies examined participants with COPD only. Seventeen studies reported baseline lung function values; most indicated a moderate‐to‐severe degree of airflow limitation. Nineteen studies indicated inclusion of participants with a mean baseline exacerbation frequency of two or more in the preceding year.
Primary outcomes
Participants administered an immunostimulant agent for a mean duration of six months were slightly more likely to be free of exacerbations during that time (OR 1.48, 95% CI 1.15 to 1.90; 15 studies, 2961 participants; moderate‐certainty evidence), although this result was moderately heterogeneous (I2 = 53%). Based on a mean estimate of baseline risk of 52%, 11 (95% CI 7 to 29) participants required treatment for this duration for one to be exacerbation‐free.
Five studies were included in meta‐analysis for all‐cause mortality, with the pooled result suggesting that immunostimulants probably result in little to no difference in all‐cause mortality measured over a mean duration of eight months, although the CIs were not sufficiently narrow to exclude a clinically important difference (OR 0.64, 95% CI 0.37 to 1.10; 5 studies, 1558 participants; moderate‐certainty evidence).
Immunostimulants increased the odds of participants being exacerbation‐free in all but three studies (Braido 2015; Collet 1997; EUCTR2007‐004702‐27‐DE). Braido 2015 examined the effect of the Ismigen compared to placebo over 12 months in 288 participants. The study was well‐described and at low risk of bias for half of the domains, with no high‐risk judgement in any domain. Study authors found no difference between the proportion of participants with no exacerbations in the intervention (70.55%) and placebo (71.13%) groups, postulating that the reasons for this may have been due to the low general exacerbation rate seen among study participants (greater than 70% participants did not experience an exacerbation over the study duration), and 27% dropout rate across the study (well‐described, ITT analysis used). Similarly Collet 1997, which examined the effect of OM‐85 over six months in 381 participants, was at low risk of bias for half of the domains, with no high‐risk judgement in any domain. Authors reported no difference between intervention and placebo with respect to the number of participants experiencing one or more exacerbations during the study period (P = 0.872, as reported in the text of the paper). Reported proportions equated to 55.5% of participants in the immunostimulant group and 56.3% of participants in the placebo group with no exacerbations during the study period. It was suggested by the authors that this observed result may have been related to more frequent recording of respiratory events that did not meet the criteria definition for an exacerbation in the placebo compared to the immunostimulant group. In EUCTR2007‐004702‐27‐DE, where trialists analysed the effect of OM‐85 versus placebo in 357 participants over six months, the negative direction of the point estimate of effect for the immunostimulant group was more noticeable (OR 0.81, 95% CI 0.52 to 1.24), although CIs were wide and included the possibility of no difference or a positive effect. Of note, this study was at high risk for attrition bias and selective outcome reporting; the study authors, as well as an independent group of experts, considered the study to be flawed due to the degree of ineligible and missing data, and considered that conclusions could not be drawn from this result. All three studies were associated with pharmaceutical company sponsorship.
Secondary outcomes
Immunostimulants may result in little to no difference in respiratory‐related mortality when comparing the odds in intervention and placebo groups over a mean duration of six months (OR 0.40, 95% CI 0.15 to 1.07; 2 studies, 735 participants; low‐certainty evidence); however, the upper limit of the CI for the estimate also included the possibility of no difference between groups. Furthermore, certainty of this estimate was lessened by the inclusion of only two studies in meta‐analysis, one of which included a population of participants that may not be generalisable to a broader population of participants with COPD or chronic bronchitis (or both).
Immunostimulants may be associated with an improvement in HRQoL scores, measured by the SGRQ, but the evidence was very uncertain. The pooled effect estimate from two studies indicated a small increase in the odds of improvement in scores with immunostimulant compared to placebo (MD −4.59, 95% CI −7.59 to −1.59), and met the MCID score of 4 units. However, the CI upper limit did not clear this threshold, and the possibility of no difference between groups could not be excluded.
The pooled result from five studies indicated that immunostimulants likely reduced the number of participants requiring antibiotics over a mean duration of six months (OR 0.34, 95% CI 0.18 to 0.63; 5 studies, 542 participants; moderate‐certainty evidence), with a moderate positive effect seen. This result had a low‐to‐moderate degree of heterogeneity (I2 = 38%), but the direction of effect was consistent across all studies.
The combined data from 20 studies that reported adverse events indicated no difference in the odds of experiencing an adverse event with immunostimulant compared to placebo, over a mean duration of six months (OR 1.01, 95% CI 0.84 to 1.21; 3780 participants; high‐certainty evidence). The CI limits for the associated risk ratio did not cross the 25% risk reduction threshold for appreciable harm or benefit (RR 1.02, 95% CI 0.92 to 1.13). An additional seven studies reported no events in both study arms; hence, were not incorporated into meta‐analysis.
We attempted to meta‐analyse data regarding the effect of immunostimulants from 17 studies that had reported on exacerbation duration; however, conclusions were unable to be drawn due to the high level of heterogeneity (I2 = 92%) and probable data skew. Results from an effect direction plot and binomial probability test indicated that a significant proportion of studies reported a direction of effect that favoured intervention, possibly indicating that immunostimulants are efficacious in reducing exacerbation duration compared to placebo (94%, 95% CI 73% to 99%; P = 0.0002). However, the degree of uncertainty associated with this estimate remained high.
Three studies reported mean respiratory‐related hospitalisation duration, two of which demonstrated a direction of effect that favoured immunostimulant compared to placebo; effect size and direction for the other was unclear, as measures of variance were not reported and unable to be estimated. Data were not interpretable from a meta‐analysis due to high levels of heterogeneity (I2 = 83%). The impact of immunostimulant on hospitalisations in general was reported using a wide variety of metrics across several studies. Three of the larger, more‐recent trials all reported reductions in total respiratory‐related hospital days over a mean duration of 10 months with immunostimulant compared to placebo.
For several outcomes, there was moderate‐to‐significant heterogeneity across the included studies. Therefore, results should be interpreted with caution. Outcomes where there was no heterogeneity were mortality (all‐cause), mortality (respiratory‐related), quality of life, and adverse events/side effects. Subgroup analyses were performed to investigate causes of heterogeneity for the primary outcome of number of participants with no exacerbations; however, heterogeneity remained high within subgroups. Subgroups that had low or no heterogeneity detected mostly contained small numbers of studies and participants, with non‐significant associated P values for the Chi2 test for heterogeneity.
Subgrouping did not modify the treatment effect estimate for either of the two primary outcomes; however, the validity of the estimates were limited by the degree of heterogeneity and uneven covariate distribution within some subgroups. Therefore, it was difficult to draw conclusions from these subgroup analyses.
Overall completeness and applicability of evidence
Of particular relevance to earlier trials, the definitions of chronic bronchitis or COPD (or both) in several included studies were not always well‐defined and, especially for trials involving participants with chronic bronchitis alone, were based on clinical criteria only. Only 17 studies reported lung function; of these, 13 included participants with moderate‐to‐severe airflow limitation (GOLD 2022). Although efforts were made to exclude those studies that may have included participants with other respiratory illnesses, such as asthma or bronchiectasis, it was not always clear what methods trialists had used to differentiate these conditions among the enrolled participants. Smoking prevalence appeared to vary widely across studies; although, many did not distinguish between the proportions of never‐smokers and ex‐smokers. Studies that were outliers with respect to mean participant age may also indicate a significant degree of disease heterogeneity across the included study participants, although the study with the lowest mean participant age was an outlier, only assessed one of the review outcomes, and was not included in meta‐analysis. All these factors may have impacted on the combinability of data within the review, contributed to the heterogeneity across several outcomes, and likely affected the generalisability of the evidence to a wider population of participants with COPD, as defined by current criteria (GOLD 2022). However, creating more stringent population criteria for study inclusion at the outset of this review may have excluded a significant proportion of studies, particularly older studies, which included participants based on disease definitions at the time of enrolment, as opposed to on the basis of more‐recently established and current gold‐standard diagnostic criteria.
Most studies included participants with a history of respiratory exacerbations; however, many involved short‐to‐medium follow‐up periods (mean study duration 6.6 months). Especially in studies where exacerbation outcomes were measured over periods of less than six months, any positive treatment effect in the intervention group may have been potentially exaggerated by short time frames of outcome measurement; conversely, having shorter follow‐up periods may not have allowed exacerbations to manifest in either group, contributing to negative or non‐significant results. It is similarly difficult to extrapolate evidence around mortality to a wider population when this outcome was measured over a relatively short mean duration of 8.4 months.
The applicability of the results of the earlier studies is likely to be affected by a degree of publication bias; it is probable that small, negative, or neutral immunostimulant trials were unpublished based on the results of a funnel plot performed for the primary outcome (Figure 4). The included smaller, positive trials were associated with much broader CIs and were given less weight in meta‐analysis; their exclusion did not appear to have a significant impact on the pooled effect estimate for the primary outcome. The effect of advances in COPD management over the years also influences interpretation of the results of the earlier, positive studies. It is unlikely that participants in these studies were exposed to what would currently be considered 'standard' non‐pharmacological and pharmacological management for COPD or chronic bronchitis (or both) (GOLD 2022).
The 36 included studies reported a wide range of outcomes that were direct or surrogate markers for exacerbation frequency or severity (or both). The outcome measures chosen for this review were restricted to those that were considered to enhance combinability of the data, and we preferentially used dichotomous measures for parametric analysis as they were deemed to be more statistically robust. Some other frequently reported outcomes in the included studies included count data presented as a continuous outcome, such as 'mean number of exacerbations'; these were not included in this review due to the likelihood of skew associated with this metric. Therefore, there be a range of similar or related outcome measures reported in the included studies that were not incorporated in the review analyses but which may be of clinical relevance.
Quality of the evidence
Overall, we graded the certainty of the evidence to be moderate, meaning that the true effect is probably close to the estimate of the effect, but that further research is likely to have an important impact on our confidence in the estimate and may change the estimate (see summary of findings Table 1). Several considerations led to downgrades in the level of certainty in the effect estimate for most main outcomes. For individual outcomes, the level of certainty varied from very low (quality of life) to high (adverse events/side effects).
One outcome was downgraded for risk of bias (quality of life); the meta‐analysis only included two studies, one of which was at high risk for attrition bias and for which standard deviation data had also been imputed. Most included studies for other outcomes were at low or unclear risk of bias for each domain, and exclusion of the studies at high risk of bias by sensitivity analysis did not impact on the treatment effect estimate. Two outcomes were downgraded for inconsistency (participants with no exacerbations and participants requiring antibiotics), due to moderate levels of heterogeneity. Three outcomes were downgraded for imprecision (all‐cause mortality, respiratory‐related mortality, and quality of life), as the CIs were not sufficiently narrow to exclude the possibility of no or little effect. One outcome was downgraded for indirectness (respiratory‐related mortality); only two studies were included in this analysis, one of which examined a group of participants that were not considered to be representative of a broader population with COPD or chronic bronchitis (or both), with this characteristic being relevant to the measured outcome. One outcome was downgraded for publication bias (quality of life), as only two studies were included in the analysis produced by the same author groups; one of which was an abstract. It was considered that there was not a sufficiently large body of evidence included in the analysis to affirm the effect estimate as a true result. Although the funnel plot indicated a likely degree of likely publication bias for the outcome of participants with no exacerbations (Figure 4), exclusion of the smaller, positive studies in a sensitivity analysis did not significantly alter the effect estimate, therefore this outcome was not downgraded.
The quality of the data for the outcome of exacerbation duration was very low (not meta‐analysed or included in the summary of findings table). Results from pooled analysis were associated with a high level of heterogeneity, which was interpreted to be due to methodological and clinical diversity related to study interventions, participant characteristics and study design, potentially variable outcome definitions and measures of effect, and the use of a continuous variable. It was not always clear whether trialists reported the mean duration of individual exacerbation episodes, or whether total cumulative exacerbation days during the study period had been averaged over the number of participants. Additionally, imputation of data was required for several included studies due to measures of variance not being reported. Interpretation of data related to mean duration of respiratory‐related hospitalisations was also affected by heterogeneity between included studies in a pooled analysis and a limited number of studies reporting this outcome.
Potential biases in the review process
We attempted to minimise bias during the review process by completing a comprehensive electronic search of all published and unpublished data. When screening titles and abstracts, we used a conservative approach for inclusion, especially considering the number of studies published in non‐English languages that could not be adequately screened at the first stage without the assistance of a translator. Two review authors independently reviewed full‐text articles and extracted data using Covidence software (Covidence) to ensure standardised data extraction methods. Where studies required translation, and translators were able to confirm that the studies met initial screening criteria, we sent standardised data extraction documents to translators to complete. Where possible, we used two independent translators per study to minimise the risk of bias.
Not all included studies were clear regarding definitions of chronic bronchitis, COPD, or what constituted a respiratory exacerbation. We accepted the trialists' definitions of these when marking studies for inclusion, given the variability in reporting across studies. Therefore, there is the possibility that some included studies contained participants or had exacerbation criteria that were not generalisable or consistent compared to other included trials, which may have impacted on the validity and applicability of the review results.
Several study analyses imputed standard deviations. Where this occurred, we used conservative estimates of P values and performed calculations in accordance with accepted practices; however, assumptions may have impacted on CI overlap and potentially increased heterogeneity. In one study, it was unclear whether reported measures of variance were standard deviations or standard errors (Hutas 1994). We contacted the study author for clarification, but this was unable to be differentiated. Based on use of standard deviation for continuous variables elsewhere in the study, we recorded values as such. Several studies did not clearly specify the number of participants analysed at the point of relevant outcome measurements (i.e. whether the number was less than at baseline due to dropouts that had not been reported); in these cases, we used an ITT approach for assuming the number of participants at any time point of the study.
Several studies reported data at discrete time intervals. Where these time intervals met our criteria of outcomes being measured over at least 12 weeks, we included the interval from baseline (as opposed to later intervals) in meta‐analyses; the rationale was that we considered participants analysed during the time interval of longest follow‐up would no longer be truly 'randomised' and representative of a baseline sample of patients, and treatment effects might have been confounded by prior exposure to the medication and study dropouts during the earlier parts of the study. Treatment of the data in this way may have influenced the combinability of these studies with others that had instead analysed end‐of‐study outcomes over longer durations, which may have contributed to heterogeneity when meta‐analysing data.
Our primary and secondary outcomes were changed following the data extraction process. This deviation from protocol was carefully considered and occurred following consensus discussion, prior to any data synthesis or analysis occurring. While the eventual outcomes were broadly congruent with our original outcomes, the ways in which these were measured were altered somewhat to suit the availability and patterns of data seen across studies; this may have introduced bias, especially if there was a degree of selective reporting of positive outcomes across studies.
Additionally, subgroup analyses required modification following data extraction to investigate for anticipated heterogeneity related to differences in treatment and study duration, which varied broadly across included studies. Again, these changes occurred prior to data synthesis or analysis occurring. We performed a post hoc subgroup analysis involving decade of study publication following data synthesis to further investigate heterogeneity that remained unexplained by the preplanned subgroupings. Therefore, the results of this analysis should be interpreted with caution.
Agreements and disagreements with other studies or reviews
The 2022 GOLD guidelines have specifically mentioned use of immunostimulant or immunoregulatory agents for preventing exacerbations in adults with COPD (GOLD 2022). However, their widespread and routine use has been limited due to a shortage of high‐quality data regarding their efficacy, and a lack of understanding of their mechanisms of action and long‐term safety profiles. The 2022 GOLD guidelines suggest that further studies are needed to examine the effects of these agents in people who are receiving current gold‐standard COPD maintenance therapy.
One previous systematic review found that the bacterial extracts O‐85, LW‐50020, and SL‐04 were associated with improved symptoms in people with COPD, chronic bronchitis, or both, and the meta‐analysis suggested a lessened exacerbation duration (MD −2.7, 95% CI −3.5 to −1.8). However, there was no difference between the extracts and placebo in preventing exacerbations (Steurer‐Stey 2004). Another systematic review looked at the efficacy of OM‐85 in preventing exacerbations in people with COPD or chronic bronchitis (or both). For the end point of one or more exacerbations, there was a non‐significant trend in favour of OM‐85 (RR 0.83, 95% CI 0.65 to 1.05); however, there were varied results across a range of other clinically important outcomes including hospitalisations and antibiotic use (Sprenkle 2005). In 2015, one meta‐analysis and systematic review examined the effects of OM‐85 in people with COPD on exacerbation rate, in addition to several other secondary clinical end points including hospitalisation duration and antibiotic use. OM‐85 was associated with a 20% reduction in exacerbation rate (RR 0.80, 95% CI 0.65 to 0.97; P = 0.03, as reported in the text of the paper) and a 39% reduction in the incidence rate of participants using antibiotics compared with placebo (RR 0.61, 95% CI 0.48 to 0.77; reported P < 0.0001) (Pan 2015). However, the authors concluded that there was not enough evidence to support the routine use of OM‐85 in people with COPD, suggesting that further larger‐scale trials needed to be undertaken.
The findings of this review suggest that immunostimulants may be associated with a small increase in the odds of remaining exacerbation‐free, which is consistent with the direction of effect for exacerbation outcomes seen in Sprenkle 2005 and Pan 2015. The finding that immunostimulant was associated with a reduced proportion of participants requiring antibiotics compared to placebo is also congruent with the findings in Pan 2015. Although our review found a positive effect direction with respect the outcomes of exacerbation duration and respiratory‐related hospitalisation duration, data were unable to be meaningfully combined in meta‐analysis and concerns around data quality and heterogeneity impacted on our certainty of these results.
Effect direction plot and sign test for the outcome of exacerbation duration.
Comparison 1: Immunostimulant versus placebo, Outcome 1: Number of participants with no exacerbations during the study period
Comparison 1: Immunostimulant versus placebo, Outcome 2: Participants with no exacerbations, by immunostimulant type
Comparison 1: Immunostimulant versus placebo, Outcome 3: Participants with no exacerbations, by baseline FEV1
Comparison 1: Immunostimulant versus placebo, Outcome 4: Participants with no exacerbations, by baseline exacerbation frequency
Comparison 1: Immunostimulant versus placebo, Outcome 5: Participants with no exacerbations, by treatment duration
Comparison 1: Immunostimulant versus placebo, Outcome 6: Participants with no exacerbations, by total study duration
Comparison 1: Immunostimulant versus placebo, Outcome 7: Participants with no exacerbations, by decade of publication
Comparison 1: Immunostimulant versus placebo, Outcome 8: Mortality (all‐cause)
Comparison 1: Immunostimulant versus placebo, Outcome 9: Mortality (all‐cause), by immunostimulant type
Comparison 1: Immunostimulant versus placebo, Outcome 10: Mortality (all‐cause), by baseline FEV1
Comparison 1: Immunostimulant versus placebo, Outcome 11: Mortality (all‐cause), by baseline exacerbation frequency
Comparison 1: Immunostimulant versus placebo, Outcome 12: Mortality (all‐cause), by treatment duration
Comparison 1: Immunostimulant versus placebo, Outcome 13: Mortality (all‐cause), by total study duration
Comparison 1: Immunostimulant versus placebo, Outcome 14: Mortality (respiratory‐related)
Comparison 1: Immunostimulant versus placebo, Outcome 15: Quality of life (total score St George's Respiratory Questionnaire)
Comparison 1: Immunostimulant versus placebo, Outcome 16: Number of participants requiring antibiotics
Comparison 1: Immunostimulant versus placebo, Outcome 17: Exacerbation duration
Comparison 1: Immunostimulant versus placebo, Outcome 18: Hospitalisation duration (respiratory‐related)
Comparison 1: Immunostimulant versus placebo, Outcome 19: Adverse events/side effects
| Patient or population: adults with chronic bronchitis or chronic obstructive pulmonary disease | ||||||
| Outcomes | Anticipated absolute effects* (95% CI) | Relative effect | № of participants | Certainty of the evidence | Comments | |
|---|---|---|---|---|---|---|
| Risk with placebo | Risk with immunostimulant | |||||
| Number of participants with no exacerbations during the study period | Low | OR 1.48 | 2961 | ⊕⊕⊕⊝ | Immunostimulants likely result in a slight increase in the number of participants with no exacerbations (or inversely, result in a reduction in the number of participants with ≥ 1 exacerbations). | |
| 5 per 100a | 7 per 100 | |||||
| High | ||||||
| 68 per 100a | 76 per 100 | |||||
| Mortality (all‐cause) | Low | OR 0.64 | 1558 | ⊕⊕⊕⊝ | Immunostimulants probably result in little to no difference in all‐cause mortality. | |
| 21 per 1000a | 14 per 1000 | |||||
| High | ||||||
| 58 per 1000a | 38 per 1000 | |||||
| Mortality (respiratory‐related) | 4 per 100 | 2 per 100 | OR 0.40 | 735 | ⊕⊕⊝⊝ | Immunostimulants may result in little to no difference in respiratory‐related mortality. |
| Quality of life | The mean quality of life was 37.5 points | MD 4.59 points lower | ‐ | 617 | ⊕⊝⊝⊝ | Immunostimulants may be associated with improvement in health‐related quality‐of‐life scores, but the evidence is very uncertain. |
| Number of participants requiring antibiotics | Low | OR 0.34 | 542 | ⊕⊕⊕⊝ | Immunostimulants likely result in a reduction in the number of participants requiring antibiotics. | |
| 60 per 100a | 34 per 100 | |||||
| High | ||||||
| 73 per 100a | 48 per 100 | |||||
| Adverse events/side effects | Low | OR 1.01 | 3780 | ⊕⊕⊕⊕ | Immunostimulants do not increase the number of participants with an adverse event. | |
| 6 per 100a | 6 per 100 | |||||
| High | ||||||
| 44 per 100a | 44 per 100 | |||||
| *The risk in the intervention group (and its 95% confidence interval) is based on the assumed risk in the comparison group and the relative effect of the intervention (and its 95% CI). | ||||||
| GRADE Working Group grades of evidence | ||||||
| See interactive version of this table: https://gdt.gradepro.org/presentations/#/isof/isof_question_revman_web_429628585692116898. | ||||||
| a The lowest and highest risk values are the second‐lowest and second‐highest proportions of participants with no exacerbations in the control groups from the studies included in this review. | ||||||
| Study ID | Total na | Study duration (weeks) | Mean age (years) | Participant type | Presence of acute exacerbation as an inclusion criteria | Intervention (total duration) | Category | Route | Relevant outcomes measured |
|---|---|---|---|---|---|---|---|---|---|
| 344 | 26 | 67.7 | COPD | No | AM3 1 g 3 times daily (6 months) | Candida utilis polysaccharide/protein compound | Oral | Participants with/without an exacerbation, number of exacerbations, SGRQ score, AEs | |
| 364 | 13 | 57.7 | COPD | No | AM3 3 g daily (3 months) | Candida utilis polysaccharide/protein compound | Oral | SGRQ score | |
| 110 | 26 | 62.9 | CB/COPD (data extracted limited to COPD patient subset) | No | RU41740 (Biostim) 2 mg daily for 8 consecutive days in first month; 1 mg daily for 8 consecutive days in second/third months (3 months) | Bacteria‐derived | Oral | Participants with/without an exacerbation, number of exacerbations, exacerbation duration, number of participants requiring antibiotics, AEs | |
| 181 | 17 | 62.3 | CB | No | Pidotimod 800 mg daily (2 months) | Synthetic agent | Oral | Participants with/without an exacerbation, AEs | |
| 184 | 52 | 69.2 | COPD (asthma patient subset excluded from analysis) | No | D53 (Ribomunyl) 4 sequences of 15 days of aerosol treatment separated by 1‐week intervals. Subcutaneous injections days 7 and 14 of first sequence and day 14 of following sequences (2.7 months) | Bacteria‐derived | Aerosol and subcutaneous | Mean number of exacerbations per participant, exacerbation duration, participants with no or a reduction in antibiotic therapy, AEs | |
| 172 | 26 | 60.5 | CB/COPD | No | RU41740 (Biostim) 2 mg daily or 8 mg daily (2 intervention groups) for 1 week, alternate weeks (3 months) | Bacteria‐derived | Oral | Participants with/without an exacerbation, exacerbation duration, duration of antibiotic therapy, AEs, mortality (all‐cause) | |
| 40 | 17 | 70.0 | CB/COPD | No | AM3 500 mg 3 times daily (4 months) | Candida utilis polysaccharide/protein compound | Oral | Participants with/without an exacerbation, mean number of exacerbations per participant, number of exacerbations, exacerbation duration, AEs | |
| 288 | 52 | 69.0 | COPD | No | Ismigen 50 mg daily for 10 consecutive days/month for 3 months, then 3 months without treatment, then repeat of the initial regimen (9 months) | Polyvalent mechanical bacterial lysate | Sublingual | Participants with/without an exacerbation, exacerbation rate, days to first exacerbation, hospitalisation days (respiratory and all‐cause), participants requiring concomitant medications, QoL scale scores, AEs, mortality (all‐cause) | |
| 40 | 13 | 65.0 | COPD | No | AM3 500 mg 3 times daily (3 months) | Candida utilis polysaccharide/protein compound | Oral | Mean number of exacerbations per participant, AEs | |
| 236 | 13 | 64.9 | COPD plus "cell‐mediated immune deficiency" | No | Thymomodulin 60 mg twice daily (3 months) | Thymic extract | Oral | Exacerbation rate, QoL scale scores, AEs | |
| 178 | 13 | 66.5 | COPD | No | Ismigen 50 mg daily for 10 consecutive days/month for 3 months (3 months) | Polyvalent mechanical bacterial lysate | Sublingual | Exacerbation rate, exacerbation duration, duration of antibiotic therapy, hospitalisation rate (respiratory), hospitalisation duration (respiratory), hospitalisation days (respiratory), AEs, mortality (all‐cause) | |
| 494 | 22 | 65.7 | CB/COPD | No | Pidotimod 800 mg daily (2 months) | Synthetic agent | Oral | Exacerbation rate, days to first exacerbation, exacerbation duration, duration of antibiotic therapy, AEs | |
| 381 | 26 | 66.1 | COPD | No | OM‐85 7 mg daily for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (4 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, mean number of exacerbations per participant, hospitalisation rate (respiratory and all‐cause), participants requiring hospitalisation (respiratory and all‐cause), hospitalisation duration (respiratory), hospitalisation days (respiratory and all‐cause), mean number of hospital days per participant (respiratory), change in SF‐36 scale scores, AEs, mortality (respiratory and all‐cause) | |
| 104 | 26 | 48.2 | CB/COPD | Yes | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Exacerbation duration, total exacerbation days, participants requiring antibiotics, participants requiring bronchodilator therapy, FEV1, AEs | |
| 265 | 26 | 81.8 | CB/COPD | Yes | OM‐85 7 mg daily for 10 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, number of exacerbations, participants requiring antibiotic therapy, AEs | |
| 60 | 17 | 62.7 | CB plus "borderline immune deficiency" | No | Lantigen B (2 intervention groups) 15 drops twice daily for 1 month, then 1 month without treatment, then a 15 days of initial regimen (2.5 months) | Bacterial lysate | Sublingual | Mean number of exacerbations per participant, exacerbation duration, participants requiring antibiotics, AEs | |
| 59 | 26 | 45.5 | CB | Yes | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Mean number of exacerbations per participant, exacerbation duration, duration of concomitant therapy, FEV1 | |
| 357 | 26 | Not reported | COPD | No | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, number of exacerbations, exacerbation duration, days to first exacerbation, duration of concomitant therapy, hospitalisation rate and duration (all‐cause), SGRQ scores, FEV1, AEs/serious AEs, mortality (all‐cause) | |
| 29 | 39 | 57.0 | CB/COPD | No | RU41740 (Biostim) 2 mg daily for 8 consecutive days in first month; 1 mg daily for 8 consecutive days in second/third months (3 months) | Bacteria‐derived | Oral | Participants with/without an exacerbation, mean number of exacerbations per participant, exacerbation duration, AEs | |
| 64 | 26 | 46.0 | CB | No | D53 (Ribomunyl) 1 tablet (dose not specified) daily for 4 consecutive days/week for 3 weeks, then 1 tablet daily for 4 consecutive days/month for 5 months (5.75 months) | Bacteria‐derived | Oral | Mean number of exacerbations per participant, exacerbation duration, participants requiring antibiotics, AEs | |
| 136 | 60 | 47.3 | CB | No | Symbioflor 30 drops 3 times daily (6 months) | Bacteria‐derived | Oral vs sublingual (liquid preparation) | Participants with/without an exacerbation, number of exacerbations, days to first exacerbation, participants requiring antibiotics, AEs | |
| 114 | 26 | 51.7 | CB/COPD | No | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Exacerbation duration, duration of antibiotic therapy | |
| 81 | 26 | 57.0 | CB/COPD | Yes | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Number of exacerbations, duration of antibiotic therapy, duration of corticosteroid therapy, hospitalisation rate (respiratory) | |
| 90 | 52 | 66.0 | CB/COPD | No | OM‐85 7 mg daily for 10 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral | Mean number of exacerbations per participant, exacerbation duration, duration of antibiotic therapy, AEs | |
| 44 | 26 | 47.7 | COPD | No | Diribiotine CK 10 mL daily for 20 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral (liquid preparation) | Participants with/without an exacerbation, mean number of exacerbations per participant, exacerbation duration, participants requiring antibiotics, AEs | |
| 79 | 26 | 55.1 | CB | Yes | OM‐85 7 mg daily for 10 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral | AEs | |
| 340 | 22 | Not reported | COPD | No | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Exacerbation rate, exacerbation duration, AEs | |
| 354 | 26 | 82.0 | CB/COPD | No | OM‐85 7 mg daily for 10 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, number of exacerbations, mean number of antibiotic courses, participants requiring bronchodilator therapy, participants requiring corticosteroids, AEs, mortality (respiratory and all‐cause) | |
| 19 | 17 | 52.8 | CB | No | OM‐85 7 mg daily for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (4 months) | Bacterial lysate | Oral | Mean number of exacerbations per participant, exacerbation duration, hospitalisation duration (respiratory), mean number of hospitalisations per participant (respiratory), AEs | |
| 88 | 52 | Not reported | CB/COPD | No | Thymomodulin 80 mg 3 times daily (3 months) | Thymic extract | Oral | Mean number of exacerbations per participant, exacerbation duration, hospitalisation duration (all‐cause), mean number of concomitant medication courses, duration of concomitant therapies, AEs | |
| 55 | 26 | 55.7 | CB | No | Diribiotine CK 10 mL daily for 20 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral (liquid preparation) | Mean number of exacerbations per participant, exacerbation duration, mean number of antibiotic courses, duration of antibiotic therapy, AEs | |
| 273 | 26 | 57.6 | CB/COPD | Yes | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, mean number of exacerbations per participant, number of exacerbations, participants requiring concomitant medication, AEs, mortality (all‐cause) | |
| 50 | 26 | 35.2 | CB | No | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Mean number of exacerbations per participant, exacerbation duration, duration of antibiotic therapy, duration of bronchodilator therapy, AEs | |
| 428 | 22 | 63.1 | CB/COPD | Yes | OM‐85 7 mg daily for 10 consecutive days/month for 3 months (3 months) | Bacterial lysate | Oral | Participants with/without an exacerbation, participants requiring concomitant medication, AEs | |
| 29 | 35 | 61.0 | CB/COPD | No | Hyaluronan 7.5 mg subcutaneously weekly (6 months) | Synthetic agent | Subcutaneous | Participants with/without an exacerbation, number of exacerbations, total exacerbation days, total antibiotic days, AEs | |
| 62 | 26 | 57.9 | COPD | No | OM‐85 7 mg daily for 1 month, no treatment for 1 month, then 7 mg daily for 10 consecutive days/month for 3 months (5 months) | Bacterial lysate | Oral | Number of exacerbations, exacerbation duration, participants requiring antibiotics, participants requiring bronchodilator therapy | |
| AE: adverse events; CB: chronic bronchitis; COPD: chronic obstructive pulmonary disease; QoL: quality of life; SF‐36: 36‐item Short Form Survey; SGRQ: St George's Respiratory Questionnaire. | |||||||||
| Trade name | Generic name | Active entity |
|---|---|---|
| Adimod | Pidotimod | Synthetic agent |
| Biostim | RU41740 | Bacteria‐derived (glycoproteins and membrane fractions of Klebsiella pneumoniae) |
| Broncho‐Vaxom, Broncho‐Munal, Ommunal, Paxoral, Vaxoral | OM‐85 | Bacterial lysate |
| Diribiotine CK | Not available | Bacterial lysate |
| Hymovis, Monovisc, Orthovisc | Hyaluronan | Synthetic agent |
| Immunoferon, Inmunol | AM3 | Glycophosphopeptical (polysaccharide and protein compounds of Candida utilis) |
| Ismigen | Not available | Polyvalent mechanical bacterial lysate |
| Lantigen B | Not available | Bacterial lysate |
| Ribomunyl, Ribovac, Immucytal | D53 | Bacteria‐derived (proteoglycans of Klebsiella pneumoniae and ribosomal fragments from a range of bacterial pathogens) |
| Thymolin, Leucotrofina | Thymomodulin | Thymic extract |
| Symbioflor | Not available | Bacteria‐derived (components of Enterococcus faecalis) |
| Outcome or subgroup title | No. of studies | No. of participants | Statistical method | Effect size |
| 1.1 Number of participants with no exacerbations during the study period Show forest plot | 15 | 2961 | Odds Ratio (IV, Random, 95% CI) | 1.48 [1.15, 1.90] |
| 1.2 Participants with no exacerbations, by immunostimulant type Show forest plot | 15 | 2961 | Odds Ratio (IV, Random, 95% CI) | 1.48 [1.15, 1.90] |
| 1.2.1 OM‐85 | 6 | 1933 | Odds Ratio (IV, Random, 95% CI) | 1.32 [0.97, 1.79] |
| 1.2.2 AM3 | 2 | 293 | Odds Ratio (IV, Random, 95% CI) | 1.35 [0.83, 2.20] |
| 1.2.3 Biostim | 3 | 253 | Odds Ratio (IV, Random, 95% CI) | 2.39 [1.00, 5.70] |
| 1.2.4 Other | 4 | 482 | Odds Ratio (IV, Random, 95% CI) | 2.59 [0.88, 7.59] |
| 1.3 Participants with no exacerbations, by baseline FEV1 Show forest plot | 8 | 2070 | Odds Ratio (IV, Random, 95% CI) | 1.34 [1.06, 1.71] |
| 1.3.1 FEV1 ≥ 50% predicted | 5 | 1264 | Odds Ratio (IV, Random, 95% CI) | 1.50 [1.03, 2.17] |
| 1.3.2 FEV1 < 50% predicted | 3 | 806 | Odds Ratio (IV, Random, 95% CI) | 1.15 [0.87, 1.52] |
| 1.4 Participants with no exacerbations, by baseline exacerbation frequency Show forest plot | 15 | 2961 | Odds Ratio (IV, Random, 95% CI) | 1.48 [1.15, 1.90] |
| 1.4.1 Mean exacerbation rate ≥ 2 in preceding year | 10 | 2064 | Odds Ratio (IV, Random, 95% CI) | 1.53 [1.13, 2.07] |
| 1.4.2 Mean exacerbation rate < 2 in preceding year or not specified | 5 | 897 | Odds Ratio (IV, Random, 95% CI) | 1.43 [0.87, 2.35] |
| 1.5 Participants with no exacerbations, by treatment duration Show forest plot | 15 | 2961 | Odds Ratio (IV, Random, 95% CI) | 1.48 [1.15, 1.90] |
| 1.5.1 Duration ≤ 3 months | 7 | 1221 | Odds Ratio (IV, Random, 95% CI) | 1.73 [1.23, 2.43] |
| 1.5.2 Duration > 3 months | 8 | 1740 | Odds Ratio (IV, Random, 95% CI) | 1.30 [0.93, 1.82] |
| 1.6 Participants with no exacerbations, by total study duration Show forest plot | 15 | 2961 | Odds Ratio (IV, Random, 95% CI) | 1.48 [1.15, 1.90] |
| 1.6.1 Duration 3 to < 6 months | 3 | 241 | Odds Ratio (IV, Random, 95% CI) | 2.39 [0.72, 7.89] |
| 1.6.2 Duration 6 to < 12 months | 11 | 2432 | Odds Ratio (IV, Random, 95% CI) | 1.52 [1.14, 2.03] |
| 1.6.3 Duration ≥ 12 months | 1 | 288 | Odds Ratio (IV, Random, 95% CI) | 0.97 [0.58, 1.62] |
| 1.7 Participants with no exacerbations, by decade of publication Show forest plot | 15 | 2961 | Odds Ratio (IV, Random, 95% CI) | 1.48 [1.15, 1.90] |
| 1.7.1 Years 2010–2019 | 2 | 672 | Odds Ratio (IV, Random, 95% CI) | 1.08 [0.79, 1.48] |
| 1.7.2 Years 2000–2009 | 4 | 1002 | Odds Ratio (IV, Random, 95% CI) | 1.39 [0.87, 2.24] |
| 1.7.3 Years 1990–1999 | 4 | 965 | Odds Ratio (IV, Random, 95% CI) | 1.64 [0.96, 2.79] |
| 1.7.4 Years 1980–1989 | 5 | 322 | Odds Ratio (IV, Random, 95% CI) | 2.74 [1.27, 5.92] |
| 1.8 Mortality (all‐cause) Show forest plot | 5 | 1558 | Odds Ratio (IV, Random, 95% CI) | 0.64 [0.37, 1.10] |
| 1.9 Mortality (all‐cause), by immunostimulant type Show forest plot | 5 | 1558 | Odds Ratio (IV, Random, 95% CI) | 0.64 [0.37, 1.10] |
| 1.9.1 OM‐85 | 3 | 1092 | Odds Ratio (IV, Random, 95% CI) | 0.69 [0.38, 1.26] |
| 1.9.2 Ismigen | 2 | 466 | Odds Ratio (IV, Random, 95% CI) | 0.47 [0.14, 1.60] |
| 1.10 Mortality (all‐cause), by baseline FEV1 Show forest plot | 4 | 1201 | Odds Ratio (IV, Random, 95% CI) | 0.65 [0.38, 1.13] |
| 1.10.1 FEV1 ≥ 50% predicted | 2 | 642 | Odds Ratio (IV, Random, 95% CI) | 0.79 [0.42, 1.47] |
| 1.10.2 FEV1 < 50% predicted | 2 | 559 | Odds Ratio (IV, Random, 95% CI) | 0.34 [0.11, 1.09] |
| 1.11 Mortality (all‐cause), by baseline exacerbation frequency Show forest plot | 5 | 1558 | Odds Ratio (IV, Random, 95% CI) | 0.64 [0.37, 1.10] |
| 1.11.1 Mean exacerbation rate ≥ 2 in preceding year | 3 | 889 | Odds Ratio (IV, Random, 95% CI) | 0.70 [0.38, 1.29] |
| 1.11.2 Mean exacerbation rate < 2 in preceding year or not specified | 2 | 669 | Odds Ratio (IV, Random, 95% CI) | 0.44 [0.13, 1.46] |
| 1.12 Mortality (all‐cause), by treatment duration Show forest plot | 5 | 1558 | Odds Ratio (IV, Random, 95% CI) | 0.64 [0.37, 1.10] |
| 1.12.1 Duration ≤ 3 months | 2 | 532 | Odds Ratio (IV, Random, 95% CI) | 0.72 [0.39, 1.34] |
| 1.12.2 Duration > 3 months | 3 | 1026 | Odds Ratio (IV, Random, 95% CI) | 0.42 [0.14, 1.31] |
| 1.13 Mortality (all‐cause), by total study duration Show forest plot | 5 | 1558 | Odds Ratio (IV, Random, 95% CI) | 0.64 [0.37, 1.10] |
| 1.13.1 Duration 6 to < 12 months | 3 | 1092 | Odds Ratio (IV, Random, 95% CI) | 0.69 [0.38, 1.26] |
| 1.13.2 Duration ≥ 12 months | 2 | 466 | Odds Ratio (IV, Random, 95% CI) | 0.47 [0.14, 1.60] |
| 1.14 Mortality (respiratory‐related) Show forest plot | 2 | 735 | Odds Ratio (IV, Random, 95% CI) | 0.40 [0.15, 1.07] |
| 1.15 Quality of life (total score St George's Respiratory Questionnaire) Show forest plot | 2 | 617 | Mean Difference (IV, Random, 95% CI) | ‐4.59 [‐7.59, ‐1.59] |
| 1.16 Number of participants requiring antibiotics Show forest plot | 5 | 542 | Odds Ratio (IV, Random, 95% CI) | 0.34 [0.18, 0.63] |
| 1.17 Exacerbation duration Show forest plot | 17 | Mean Difference (IV, Random, 95% CI) | Totals not selected | |
| 1.18 Hospitalisation duration (respiratory‐related) Show forest plot | 2 | Mean Difference (IV, Random, 95% CI) | Totals not selected | |
| 1.19 Adverse events/side effects Show forest plot | 20 | 3780 | Odds Ratio (IV, Random, 95% CI) | 1.01 [0.84, 1.21] |
