Rectal nonsteroidal anti-inflammatory drugs (NSAIDs) are not commonly used clinically for preventing post-endoscopic retrograde cholangiopancreatography (ERCP) pancreatitis. To evaluate the efficacy and safety of NSAIDs for post-ERCP prophylaxis, we systematically reviewed sixteen randomized controlled trials (involving 6458 patients) that compared rectal NSAIDs with placebo or no treatment for post-ERCP pancreatitis prophylaxis updated to August 2016. GRADE framework was used to assess the quality of evidence. There was “high quality” evidence that rectal NSAIDs were associated with significant reduction in the risk of overall post-ERCP pancreatitis . Subgroup analyses demonstrated that diclofenac  was probably superior to indomethacin , post-ERCP administration  was probably superior to pre-ERCP , and that mixed-risk population received more benefits  than average-risk population , but less than high-risk population . Moreover, “high quality” evidence showed that rectal NSAIDs were safe when given as a standard dose . In conclusion, this meta-analysis revealed that rectal NSAIDs are effective and safe in the prevention of post-ERCP pancreatitis in populations with all levels of risk. Acute pancreatitis is the most common complication of endoscopic retrograde cholangiopancreatography (ERCP). The reported frequency of post-ERCP pancreatitis varies between 1% and 13.3% in the unselected patients1256To date, many agents are explored extensively for pharmacological potential against post-ERCP pancreatitis, such as somatostatin1117182021222324251112272830So far, however, rectal NSAIDs for post-pancreatitis prophylaxis have not been widely recommended in routine clinical use. Recently, nine new RCTs1718192022242627We searched PubMed, EMBASE and the Cochrane Library for studies of NSAIDs in the prevention of post-ERCP pancreatitis updated to August 2016. Key words and/or medical subject heading terms were as follows:  AND (post-ERCP pancreatitis or post-endoscopic retrograde cholangiopancreatography pancreatitis or pancreatitis) see . ReferenArticles or abstracts were included if they met the following criteria: (1) the study was a prospective, randomized controlled trial (RCT); (2) compared NSAIDs with placebo or no treatment; (3) examined the efficacy and safety of rectally administered NSAIDs for prevention of post-ERCP pancreatitis; (4) original data not duplicated in another manuscript. Cohort studies, case-control studies, case reports and case series were excluded. The studies focusing on the role of NSAIDs in prevention of post-ERCP pancreatitis using any other routes, such as oral, intraduodenal, intravenous and intramuscular, other than rectal were also excluded. If more than one version of the same study was retrieved, only the most complete or the latest one was used.Data extraction was independently performed by 2 reviewers according to the prespecified selection criteria. Disagreement was resolved by discussion between the two authors. The following information from each studies was extracted: first author, publication year, study location, study design, patient characteristics, sample size, intervention approaches , indications and severity criteria. In addition the outcome data of studies, including the number of post-ERCP pancreatitis , amylase concentrations (serum amylase level 2 h or 24 h post-ERCP), the number of hyperamylasemia as well as pain, complications  and deaths were also included. Methodological quality of the included studies was evaluated by using criteria set forth by the Cochrane Collaboration tool for assessing the risk of biasThe quality of evidence was rated for each summary estimate through the GRADE frameworkI2statistic, a test used to quantify inconsistency across studies resulted from heterogeneity rather than from chance. The statistical heterogeneity was assessed by calculating the Cochran’ Q statistic with a significance level of P < 0.10 or I2 > 50%. Generally, an A I2 of 0–30% represents insignificant heterogeneity, I2 of 30–50% represents mild heterogeneity, I2 of 50–75% represents moderate heterogeneity, and I2 of more than 75% represents substantial heterogeneity. A meta-analysis of intention-to-treat data was performed using the random model Mantel-Haenszel method independent of heterogeneity, which could generate more conservative and reasonable results based on Q text or I2 statistic37P   < 0.05, except where otherwise specified.We performed the meta-analysis in accordance with the recommendations of Cochrane Collaborationet al.et al.Primary outcome was the risk of any post-ERCP pancreatitis. Secondary outcomes were the risk of mild or moderate to severe post-ERCP pancreatitis, 2 h or 24 h post-ERCP serum amylase level, the risk of hyperamylasemia as well as post-procedural pain, and adverse events associated with NSAIDs therapy. We also performed several subgroup analyses by grouping the type of NSAIDs (indomethacin or diclofenac), the timing of NSAIDs administration (pre- or post-ERCP), the risk of population , mean age of each study (≤60 or >60), pancreatic stent (yes or no). It should be noted that the risk stratification of the patients in included studies was defined based mainly on findings proposed by Masci P value of less than 0.1 was considered representative of statistically significant publication bias. Review Manager  was used for all analyses. All statistical tests were 2-sided.Publication bias was assessed by visual inspection of funnel plots, and further quantitative analysis by Begg adjusted rank correlation testThe initial search yielded 3537 relevant records of which 3419 were excluded because of duplicate data or based on the screening of titles and abstracts . The rem111217182021222324252627282943111217182022232425262728294321241217271112NSAIDs (diclofenac or indomethacin) were all administered rectally in the included studies, either pre-ERCP in 9 studies1718192123242526111227281117182122232425262728431211182123242526272817122011121822272822The risk of bias assessments across studies were described in P < 0.01), with a mild heterogeneity  , with insignificant heterogeneity . The quality of evidence of the efficacy of rectal NSAIDs in preventing any post-ERCP pancreatitis was rated “high quality” according to the GRADE framework. No evidence supporting publication bias was identified based on visual inspection of the funnel plot . Post-ERCP pancreatitis was documented in 170 of 3226 patients (5.0%) with rectal NSAIDs, compared with 317 of 3232 (9.9%) with no treatment. Pooling showed that rectal NSAIDs were associated with a significant reduction in overall risk of post-ERCP pancreatitis compared with patients with no treatment  . The ARRnel plot , Begg teP < 0.01 and RR = 0.41; 95% CI, 0.19–0.90; P = 0.03, respectively). Pooling estimates showed that rectal diclofenac appeared to be more effective than rectal indomethacin in preventing post-ERCP pancreatitis. Heterogeneity in indomethacin group was insignificant , however, heterogeneity in the former group was moderate . After removal of the source of heterogeneity,P < 0.01) without heterogeneity . When stratifying subgroups by timing of administration, rectal NSAIDs administered post-ERCP  were more effective than those administered pre-ERCP . Heterogeneity was moderate in former subgroup . After exclusion of the source of heterogeneityP < 0.01), with insignificant heterogeneity .Subgroup analyses were performed to explore the source of heterogeneity among studies based on key study characteristics and clinical factors. The prophylactic benefit of rectal NSAIDs (indomethacin or diclofenac) on reducing the risk of post-ERCP pancreatitis was consistently found in all of the subgroup analyses. When stratifying studies by drug type, both indomethacin and diclofenac showed a significant efficacy . More benefits could be observed in mixed-risk population  compared with average-risk population . Moderate heterogeneity  and mild heterogeneity  separately existed in high-risk and mixed-risk subgroup, although no significant heterogeneity was in average-risk subgroup . When removing the source of heterogeneityP < 0.01 or RR = 0.46; 95% CI, 0.33–0.66, P < 0.01). As younger age was a definite risk factor for post-ERCP pancreatitisP < 0.01) compared with patients with a mean age > 60 years . However, heterogeneity in latter subgroup was moderate . After removing the source of heterogeneityAfter stratification according to different risk population, it was noted that rectal NSAIDs were most effective in high-risk population . The quality of evidence for the two outcomes was rated as “high quality” and “low quality”, respectively. Six studies1923242627P = 0.03) . Two studies23P = 0.01) . The quality of evidence (GRADE) for these two outcomes was both rated as “moderate quality”.Only 5 studies122527291226.03) see . Substan.01) see , withoutP < 0.01) (P < 0.01) . The quality of evidence for this two outcomes was rated as “high quality”.Fourteen studies11171819202122232425262728111819202122232425262728 < 0.01) , and mod < 0.01) . No signP = 0.94) with no evident heterogeneity . Two studies2424P = 0.30 and RR = 0.74; 95% CI, 0.21–2.63, P = 0.64, respectively) with no evident heterogeneity . One study1922P = 0.26), with no heterogeneity . Generally, an arithmetical trend was noted that rectal NSAIDs may result in a decrease in risk of overall adverse events , with no evident heterogeneity  (Six studies1122242627 = 0.36)  Fig. 6)P = 0.94)In this meta-analysis of 16 RCTs that included 6458 patients, there was “high quality” evidence to suggest that rectal NSAIDs were associated with about a 45% decrease in the risk of post-ERCP pancreatitis. The NNT was 20. After stratifying studies according to the severity of post-ERCP pancreatitis, “high quality” evidence showed that a significant reduction was recognized in both mild and moderate to severe post-ERCP pancreatitis . Specific analyses for amylase concentration, hyperamylasemia, and pain indicated that there were “high quality”, “low quality”, “moderate quality” and “moderate quality” evidence that rectal NSAIDs could prevent the elevation of serum amylase 2 h or 24 h post-ERCP, lower the risk of hyperamylasemia and relieve the post-procedural pain, respectively. In terms of safety, we also found “high evidence” that no statistical difference in adverse events potentially attributable to rectal NSAIDs therapy was detected, strongly indicating that rectal NSAIDs are safe when given as standard doses (100 mg or 50 mg). Although mild heterogeneity existed in overall post-ERCP pancreatitis, the magnitude or direction of the summary effect remains unchanged in multiple sensitivity analyses.et al.et al.et al.et al.The present findings are consistent with the current ESGE and JSHBPS guidelines, in which NSAIDs (diclofenac or indomethacin) are widely recommended to administer rectally either pre- or post-endoscopic procedure in preventing post-ERCP pancreatitis for all risk patients30et al.1447et al.In the present meta-analysis, we explore whether the drug type or administrated timing of rectal NSAIDs had an impact on the outcome. On a subgroup analysis in which the type of NSAIDs (diclofenac or indomethacin) was compared for prevention of pancreatitis, we found that rectal diclofenac seemed more effective than rectal indomethacin for prevention of post-ERCP pancreatitis. As the discovered pharmacological mechanism in animal model, inactivation of phospholipase A2 could result in the inhibition of several important classes of pro-inflammatory lipids, e.g., prostaglandins, leukotrienes and platelet activating factor. In the case of phospholipase A2 inhibition, however, indomethacin demonstrated a stronger inhibitory action compared with diclofenac. As statistics go, rectal diclofenac appeared to be more effective than rectal indomethacin in preventing post-ERCP pancreatitis. However, it should be noted that the number of subjects included in diclofenac group amounted to roughly a seventh of indomethacin group. The confidence interval (0.19–0.90) was quite wide duo to the fact that only 915 subjects were included in 6 studies as opposed to 5543 subjects of indomethacin group in 10 studies, and this should only be considered a very rough estimate of the comparison of rectal diclofenac with rectal indomethacin. To date, there was no larger RCT to perform to compare rectal indomethacin with diclofenac in a head-to-head form to explore any difference between these 2 agents regarding the efficacy in prevention of post-ERCP pancreatitis. Our aforementioned findings differ from the meta-analysis reported by Ding et al.et al.et al.et al. pooled 8 RCTs for their meta-analysis, while the present meta-analysis included a total of 16 RCTs. These extra studies contributed 40.2% of the total random effects weight in the present meta-analysis. Furthermore, Inamdar et al. did not assess safety of the drug. In our subgroup analyses by grouping high risk, mixed risk and average risk, which suggested that rectal NSAIDs exhibit more protective benefits in lowering the risk of post-ERCP pancreatitis for high-risk population  compared with mixed- and average- risk population . To some extent, however, results of present meta-analysis may more accurately reflect the real effects of rectal NSAIDs in preventing post-ERCP pancreatitis for patients categorized by different patient- or procedure-related risk factors due to wider evidence base and sufficient data with greater statistical power. In addition, we did a subgroup analysis on the basis of age, in which mean age 60 in treatment group was deemed to be the cutoff point. The subgroup analysis reveals that rectal NSAIDs may be useful for the reduction of post-ERCP pancreatitis in younger patients (<60) compared with patients over the age of 60. The study reported by Sotoudehmanesh et al.et al.23261924et al.In the past 10 years, numerous meta-analyses have been published to investigate the preventive effect of NSAIDs on post-ERCP pancreatitis; nearly 10 meta-analyses dedicated to assessing efficiency of rectal route on prevention of post-ERCP pancreatitis. Of these meta-analyses, the earlier 2 meta-analyses by Sun, H.L. Limitations of this meta-analysis must be considered. First, we found mild heterogeneity across the studies in our meta-analysis. It is not surprising given that the difference in the data source, study population, NSAIDs type, the timing of administration, and study design. However, after exclusion of heterogeneity resource, the overall estimates of prophylactic efficacy do not change substantially. Second, the characteristics of included patients, diagnostic criterion of pancreatitis as well as criterion of pancreatitis severity, definition of the risk stratification of the patients and intervention regiment varied across studies, which may influence the results, thereby limiting comparability to some extent. However, pooled results remained robust in multiple sensitivity analyses. Third, we analyzed amylase concentration (24 h post-ERCP), the risk of hyperamylasemia and pain on the basis of insufficiency sample size, of which evidence was ranked as moderate or low based on GRADE framework. Significant differences between two groups might result from type 1 error, and need to be further investigated. Ideally, large-scale RCTs should be conducted in the future to compare different NSAIDs (indomethacin versus diclofenac), different doses and timing of administration (pre-procedure versus post-procedure), to determine the best NSAID, optimal dose and timing of administration.In conclusion, this meta-analysis provides “high quality” evidence that rectal NSAIDs appear to be a safe, simple, economical strategy, and could significantly reduce the risk of post-ERCP pancreatitis in all levels of risk populations. Therefore, rectal NSAIDs should be recommended as routine clinical use for patients who undergo ERCP without exact contraindication to NSAIDs therapy or ERCP.How to cite this article: Hou, Y.-C. et al. Efficacy and safety of rectal nonsteroidal anti-inflammatory drugs for prophylaxis against post-ERCP pancreatitis: a systematic review and meta-analysis. Sci. Rep.7, 46650; doi: 10.1038/srep46650 (2017).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


Several case-control studies reported the relationship between single nucleotide polymorphisms (SNPs) in HSP70 genes and noise-induced hearing loss (NIHL). However, their conclusions are conflicting. This meta-analysis aims to identify the association of HSP70 variants and NIHL susceptibility.2 value in the heterogeneity test.A systematical literature search was performed in PubMed, Web of Science, EMBASE, and Wanfang Chinese database. The pooled odds radio (OR), 95% confidence interval (CI) and p value were calculated in fixed- or random-effects model according to the IFour articles containing five studies, including 633 cases and 926 controls, were included. Under the allele, homozygote and dominant model, the pooled ORs  of rs1061581 were 1.32 , 1.93  and 1.455 , respectively. In addition, a significant association was found between rs2227956 in Caucasians and the NIHL susceptibility under all five genetic models. We did not discover evidence sufficient to prove the associations between the other three SNPs  and the NIHL susceptibility.This meta-analysis indicated that the two HSP70 variants, rs1061581 and rs2227956, may serve as genetic susceptibility factors for NIHL. Larger scale studies are required to further update the results. Noise is one of the most common sources of environmental stress in our contemporary society . ContinuSOD2 SOD2 [GSTnd MYH14 , but theRecent studies demonstrated that the disturbance of cellular proteostasis is critical in the development of NIHL . The heaThere is no meta-analysis or genome-wide association studies (GWAS) on the association of HSP70 polymorphisms with NIHL susceptibility to date. Aiming at evaluating the potential value of HSP70 variants in prediction for individual NIHL susceptibility, we focus on five reported SNPs in HSP70 genes here, perform a meta-analysis to address these conflicting results and assess whether HSP70 polymorphisms are associated with the susceptibility to NIHL. The results of this meta-analysis will provide theoretical basis for the application of the SNPs in HSP70 genes in the individualized prevention system to NIHL.th, 2017. We also manually checked all articles listed in the reference lists of the retrieved literatures.A comprehensive literature search was performed in the following English and Chinese databases: (1) PubMed; (2) Web of Science; (3) EMBASE; and (4) Wanfang Chinese database. The MeSH and free terms were all included in our search terms, which are listed as follows: “heat shock protein 70”, “hsp70”, “noise”, “hearing loss”, “noise-induced hearing loss” and “NIHL”. Our search logic in the PubMed database is listed as follows: “((heat shock protein 70) OR (hsp70)) AND ((noise AND (hearing loss) OR (noise-induced hearing loss) OR  OR NIHL)”. The publication languages were limited to English and Chinese. All studies that we searched were published before April 20All studies included in our meta-analysis need to be confirmed with the following criteria: (1) independent case-control studies investigating the relationship between the SNPs in HSP70 genes and the development of NIHL; (2) studies including sufficient and definite original data (the genotype frequency of each SNP in HSP70 genes in the case and control groups) in which the odds ratio (OR) with its 95% confidential interval (CI) of each genotype at every SNP site can be calculated; (3) two independent sample sets in one study were considered as two different studies; and (4) the data in the latest publication were used when duplicate publications were found.2 value and D’ value) between SNPs in different populations was extracted from the SNAP website (http://www.broadinstitute.org/mpg/snap) based on the data from the 1000 Genome Project.Data in the included studies were independently extracted by three authors  with the same “Data Extraction Form”. The following information was extracted from every included study: the mutation site of each SNP, first author’s last name, publication year, country, ethnicity, workplace of each sample set, standard of noise exposure, diagnosis criteria of hearing impairment or NIHL susceptibility, numbers of cases and controls, and consistency with Hardy-Weinberg Equilibrium (HWE) in the control group. The linkage disequilibrium (LD) pattern  was used to assess the quality of each included study. The studies with a score ≥ 7 were considered high-quality studies. When there is conflict in the process of study selection, data extraction or quality assessment, the reviewers discussed all items until they reached consensus.2 tests. The pooled ORs and 95%CIs were calculated under the fixed-effects model for p > 0.1 and I2 < 50%; otherwise, under the random-effects model with I2 ≥ 50%. Woolf’s method was applied to estimate the 95%CIs. The Z-test was applied to identify the association between the SNPs in HSP70 genes and NIHL susceptibility. We considered there was statistical significance when the overall 95% CI did not include 1 and the p-value transformed from the Z score was less than 0.05. Sub-group analysis based on ethnicity, study quality and accordance with HWE in the controls was performed. Sensitivity analysis was applied to estimate the stability of the pooled results. Publication bias for rs1043618, rs1061581 and rs2227956 were evaluated by funnel plots and the Egger’s test. For p<0.05 or a 95%CI that did not contain 0 in the Egger’s test, publication bias was considered present. HWE of the genotype distribution in the control group was evaluated by the Pearson’s X2 test. All statistical analysis was performed by Stata 13.1 software.SNPs reported in two or more studies were evaluated in a meta-analysis. The association between SNPs in the HSP70 genes and NIHL susceptibility was tested by pooled OR and 95%CI. The allele (A vs. B), homozygote (AA vs. BB), heterozygote (AA vs. AB), dominant (AA vs. AB + BB) and recessive (AA + AB vs. BB) model were all applied for genotype comparison to minimize the possibility of the second type of error. The heterogeneity among the included studies was evaluated by the Q and IOur literature selection process is presented in HSPA1B gene and NIHL susceptibility. For rs2227956, the pooled ORs and 95% CIs in Caucasian were 0.535 , 0.135 , 0.585 , 0.531  and 0.157  in the allele, homozygote, heterozygote, dominant and recessive models, respectively  in HSP70 genes for the meta-analysis. Five genetic models were used for the genotype comparison. For SNP rs1061581, two studies on Caucasian populations and one study on Asian populations containing 301 cases and 315 controls were included. The pooled ORs and 95% CIs were 1.32 , 1.93  and 1.455  in the allele, homozygous and dominant model, respectively . These rectively . For rs1We estimated the stability of the pooled ORs of the SNPs  by eliminating each of the included studies in turn. For the three SNPs mentioned above, no significant alteration of the pooled ORs was found as any of the included studies was omitted . These rThe funnel plots looked symmetric for three SNPs  in all five genetic models. We did not test the publication bias for rs2075800 and rs2763979 because of the limited number of included studies. Additionally, we did not found any publication bias via Egger’s test .NIHL affects the life quality of millions of people worldwide , 18, espG allele is a potential risk factor to NIHL susceptibility. Rs1061581 has been considered a major and synonymous SNP in HSPA1B gene, which is located at chromosome 6p21.3. Its protein product, HSP-2, is 641 amino acids in length and stress-inducible [HSPA1A is also located at chromosome 6p21.3. A variation from A to G changes methionine to threonine resulting in the 3D structure alteration of HSP-1, which may affects its function as a molecular chaperone [A significant relationship was seen between rs1061581 and NIHL susceptibility , which inducible . Howeverhaperone . We founHeterogeneity is a major problem that affects the reliability of the pooled ORs in meta-analysis. In our meta-analysis, mild to moderate heterogeneity was observed in most SNPs . The hetThere are several limitations in this meta-analysis. (1) We only searched the articles published in English and Chinese. The articles in other languages were not included in this meta-analysis, which may increase the publication bias. (2) The number of included studies is relatively low. For rs2075800 and rs2763979, only two studies were included in this meta-analysis. Their funnel plots were very asymmetric, which indicated there is significant publication bias. (3) The sample size in some included studies was relatively small, which may increase the heterogeneity. Although 633 cases and 926 controls are included in this meta-analysis, the conclusion must be updated further by including larger sample reports.G in rs1061581 and rs2227956 (only in Caucasians) may serve as genetic susceptibility factors for NIHL. However, more studies with larger sample sets from different ethnicities are needed to further confirm the relationship between HSP70 polymorphisms and NIHL susceptibility due to the limitations in this meta-analysis.To the best of our knowledge, this is the first meta-analysis on the association between the SNPs in HSP70 genes and NIHL susceptibility. Our results indicated that the allele S1 Diagram(DOC)Click here for additional data file.S1 Appendix(DOCX)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S1 Checklist(DOC)Click here for additional data file.S2 Checklist(DOCX)Click here for additional data file.S1 Fig(TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.


Meta-analysis is a statistical procedure for analyzing the combined data from different studies, and can be a major source of concise up-to-date information. The overall conclusions of a meta-analysis, however, depend heavily on the quality of the meta-analytic process, and an appropriate evaluation of the quality of meta-analysis  can be challenging. We outline ten questions biologists can ask to critically appraise a meta-analysis. These questions could also act as simple and accessible guidelines for the authors of meta-analyses. We focus on meta-analyses using non-human species, which we term ‘biological’ meta-analysis. Our ten questions are aimed at enabling a biologist to evaluate whether a biological meta-analysis embodies ‘mega-enlightenment’, a ‘mega-mistake’, or something in between. Last year saw 40 years since the coining of the term ‘meta-analysis’ by Gene Glass in 1976 . Meta-anIt is likely that the majority of us (biologists) have never conducted a meta-analysis. Chances are, however, that almost all of us have read at least one. Meta-analysis can not only provide quantitative information , but also qualitative information (such as dominant research trends and current knowledge gaps). In contrast to that of many medical and social scientists , 5, the The aim of this review is to provide ten appraisal questions that one should ask when reading a meta-analysis cf., , 19), al, al19]),When we read a biological meta-analysis, it used to be (and probably still is) common to see a statement like “a comprehensive search of the literature was conducted” without mention of the date and type of databases the authors searched. Documentation on keyword strings and inclusion criteria is often also very poor, making replication of search outcomes difficult or impossible. Superficial documentation also makes it hard to tell whether the search really was comprehensive, and, more importantly, systematic.Web of Science and Scopus (two of the most popular academic databases) is only 40–50% in many major fields [A comprehensive search attempts to identify  all relevant studies/data for a given meta-analysis, and would thus not only include multiple major databases for finding published studies, but also make use of various lesser-known databases to locate reports and unpublished studies. Despite the common belief that search results should be similar among major databases, overlaps can sometimes be only moderate. For example, overlap in search results between r fields . As wellr fields .For most meta-analyses in medicine and social sciences, the search steps are systematic and well documented for reproducibility. This is because these studies follow a protocol named the PRISMA  statement , 26; notd, Cohen’s d, Hedges’ d or Hedges’ g) and the natural logarithm (log) of the response ratio (denoted as either lnR or lnRR [z-transformed correlation coefficient (often denoted as Zr); and 3) the natural logarithm of the odds ratio (lnOR) and relative risk .A meta-analysis should not just be descriptive. The best meta-analyses ask questions or test hypotheses, as is the case with original research. The meta-analytic questions and hypotheses addressed will generally determine the types of effect size statistics the authors use –32, as w or lnRR ); 2) FisHR) for comparing survival curves [CVR) for comparing differences between the variances, rather than means, of two groups [Zr. Most biological meta-analyses will use one of the standardized effect sizes mentioned above. These effect sizes are referred to as standardized because they are unit-less (dimension-less), and thus are comparable across studies, even if those studies use different units for reporting . However, unstandardized effect sizes (raw mean difference or regression coefficients) can be used, as happens in medical and social sciences, when all studies use common and directly comparable units .We have also used and developed methods associated with less common effect size statistics such as log hazard ratio  studies. In some cases different effect sizes might be calculated for different studies in a meta-analysis and then be converted to a common type prior to analysis: for example, Zr and SMD  are inter-convertible. Thus, if we were, for example, interested in the effect of temperature on growth, we could combine results from experimental studies that compare mean growth at two temperatures (SMD) with results from observational studies that compare growth across a temperature gradient (Zr) in a single meta-analysis by transforming SMD from experimental studies to Zr [That being said, a biological meta-analysis will often bring together original studies of different types . As a general rule, SMD is considered a better fit for experimental studies, whereas es to Zr –32.Statistical non-independence occurs when data points  are somewhat related to each other. For example, multiple effect sizes may be taken from a single study, making such effect sizes correlated. Failing to account for non-independence among effect sizes (or data points) can lead to erroneous conclusions , 41–44—tmetafor [To help the reader to detect non-independence where the authors have failed to take it into account, we have illustrated four common types of dependent effect sizes in Fig. metafor .Fig. 3.CMetawin [Where non-independence goes uncorrected because of the difficulty of appropriately accounting for the correlated structure, it is usually because the non-independence is incompatible with the two traditional meta-analytic models (the fixed-effect and the random-effects models—see Q4) that are implemented in widely used software . TherefMetawin ), but maMetawin , 49. TheMetawin .vi, which is related to the sample size for each effect size; Fig. τ2, and to estimate this variance correctly requires a sample size of perhaps over ten effect sizes. Thus, random-effects models may not be appropriate for a meta-analysis with very few effect sizes, and fixed-effect models may be appropriate in such situations (bearing in mind the aforementioned assumptions). Multilevel models relax the assumptions of independence made by fixed-effect and random-effects models; that is, for example, these models allow for multiple effect sizes to come from the same study, which may be the case if one study contains several different experimental treatments, or the same experimental treatment is applied across species within one study. The simplest multilevel model depicted in Fig. There are three main kinds of meta-analytic models, which differ in their assumptions about the data being analyzed, but for all three the common and primary goal is to estimate an overall effect (but see Q5). These models are: i) fixed-effect models ; ii) rab0). The reader should also be aware that this estimated overall effect  is most commonly presented in an accompanying forest plot(s) [It is important for you, as the reader, to check whether the authors, given their data, employed an appropriate model or set of models (see Q3), because results from inappropriate models could lead to erroneous conclusions. For example, applying a fixed effect model, when a random effects model is more appropriate, may lead to errors in both the estimated magnitude of the overall effect and its uncertainty . As can  plot(s) , 56, 57.τ2, Q, and I2) are only reported in about 40% of meta-analyses [Q  is a test statistic for the between-study variance (τ2), which allows one to assess whether the estimated between-study variance is non-zero  [Q is often presented with a corresponding p value, which is interpreted in the conventional manner. However, if presented without the associated τ2, Q can be misleading because, as is the case with most statistical tests, Q is more likely to be significant when more studies are included even if τ2 is relatively small ; the reader should therefore check whether both statistics are presented. Having said that, the magnitude of the between-study variance (τ2) can be hard to interpret because it is dependent on the scale of the effect size. The heterogeneity statistic, I2, which is a type of intra-class correlation, has also been recommended as it addresses some of the issues associated with Q and τ2 [I2 ranges from 0 to 1 (or 0 to 100%) and indicates how much of the variation in effect sizes is due to the between-study variance  [The overall effect reported by a meta-analysis cannot be properly interpreted without an analysis of the heterogeneity, or inconsistency, among effect sizes. For example, an overall mean of zero can be achieved when effect sizes are all zero  or when all effect sizes are very different  but centered on zero, and clearly one should draw different conclusions in each case. Rather disturbingly, we have recently found that in ecology and evolutionary biology, tests of heterogeneity and their corresponding statistics  . As a te and τ2 , 61. I2 and 75% , 63. Howubjects) . AccordiI2 ), it is important to understand the factors, or moderators, that might explain this additional variation, because it can elucidate important processes mediating variation in the strength of effect. Moderators are equivalent to explanatory (independent) variables or predictors in a normal linear model [p hacking’; that is, non-hypothesis-based exploration for statistical significance [After quantifying variation among effect sizes beyond sampling variation Moderators can be correlated with each other  and this dependence, in turn, could lead authors to attribute an effect to the wrong moderator . For exaI2) for each subset in a subset analysis [Another way of exploring heterogeneity is to run separate meta-analysis on data subsets . This is similar to running a meta-regression with categorical moderators , with the key difference being that the authors can obtain heterogeneity statistics  can be misleading. For example, in terms of effects’ magnitudes and uncertainties, ES4 and ES6 in Fig. r = 0.1, 0.3, and 0.5, respectively [d (SMD) = 0.2, 0.5, and 0.8 [r value, but this may not be the case for SMD. Thus, it may be helpful to know that Cohen’s benchmarks for r and d are comparable. Having said that, these benchmarks, along with those for I2, have to be used carefully, because what constitute biologically important effect magnitudes can vary according to the biological questions and systems . We stress that authors should primarily be discussing their effect sizes (point estimates) and uncertainties in terms of point estimates  [Meta-analyses should focus on biological importance (which is reflected in estimated effects and their uncertainties) rather than on ectively ; for SMD and 0.8 , 61. Resls, CIs) , 70, 72.τ2) [p curve [p curve probably only works when effect sizes come from tightly controlled experiments [Meta-analysts have to assume that research is published regardless of statistical significance, and that authors have not selectively reported results  –76. Thisτ2) , 80. Somτ2) , 82 or t[p curve . The reaeriments –86 (see eriments ). A lesseriments –89. Thiseriments , 91  meta-analytic and meta-regression models , 45, 49,


P = 0.000). No evidence of publication bias was observed. Circulating high molecular weight adiponectin levels were also lower in cancer patients than in controls, with a pooled SMD of −0.502 μg/ml . This meta-analysis provides further evidence that decreased adiponectin levels is associated with risk of various cancers. Hypoadiponectinemia may represent a useful biomarker for early detection of cancers.Early detection of cancers is challenging for lack of specific biomarkers. Adiponectin is an adipokine predominantly derived from adipocytes and hypoadiponectinemia has been reported to associate with risk of many types of cancers. However, available evidence is controversial. Some studies show that increased adiponectin levels correlate with cancer risk. Therefore, we performed a meta-analysis of the association between circulating adiponectin levels and cancer development. A systematic search of PubMed, EMBASE, Wiley Online Library and Cochrane Library was conducted for eligible studies involving circulating adiponectin and malignancies from inception to August 8, 2015. Standard mean differences (SMDs) with 95% confidence intervals (95% CIs) were calculated by use of a random-effect model. Funnel plot and Egger's linear regression test were conducted to examine the risk of publication bias. 107 studies were included with 19,319 cases and 25,675 controls. The pooled analysis indicated that circulating adiponectin levels were lower in patients with various cancers than in controls, with a pooled SMD of −0.334 μg/ml (95% CI, −0.465 to −0.203,  Cancer, a major cause of human mortality, has been a worldwide public health problem. A variety of factors such as genetic lesions, environmental aspect and increasing adoption of unhealthy lifestyle are considered as crucial causes of cancer . Among tet al. in 1995, is an adipokine predominantly produced by adipocytes with the monomeric subunit containing 244 amino-acids in human and circulates abundantly in plasma [Adiponectin, firstly discovered by Scherer n plasma . Three bn plasma . Adiponen plasma , 9. Adipn plasma . A numben plasma . Besidesn plasma .It was first reported that circulating adiponectin level was lower in patients with breast cancer in 2003 . Since tThe initial comprehensive search yielded 1486 articles, of which 235 articles were excluded for duplication. Then 997 studies were ruled out because of apparent irrelevance after reading titles and/or abstracts. The remaining 254 studies were included for full-text reading, of which 151 studies were removed for one of the following reasons: (i) reviews, comments or letters (n = 37); (ii) shared population n = 13); (iii) no report of adiponectin levels and/or SDs for both patients and controls or there was not enough information to calculate them (n = 25); (iv) not case-control study (n = 76). 4 additional studies were included from checking the references list. Finally, 107 studies met the inclusion criteria and were used for further analysis [; (iii) nP values. 96 studies had NOS scores greater than 6 along with 11 studies had scores of 5. The main characteristics of eligible articles were listed in Table Among the 107 studies, a total of 25,675 controls and 19,319 cases were enrolled until August, 2015. Geographic regions were various, among which 46 studies from Asia, 39 studies from Europe, 19 studies from America, and 3 studies from Africa. 16 types of malignancies were investigated in this meta-analysis, with digestive system cancers accounting for the largest percentage (43 studies); other types included: breast cancer (20 studies), prostate cancer (13 studies), endometrial carcinoma (11 studies), lung cancer (5 studies), renal cancer (3 studies), acute leukemia (3 studies), non-Hodgkin's lymphoma (3 studies), Hodgkin's lymphoma (1 study), multiple myeloma (2 studies), melanoma (1 study), thyroid cancer (1 study), and tongue cancer (1 study). Circulating samples included serum (65 studies) and plasma (37 studies), while 5 studies did not mention the exact one. Most researches provided the mean concentrations of circulating adiponectin levels and the SDs of them. SDs from 11 studies were calculated based on the sample size and P = 0.000). Statistically significant amount of heterogeneity was observed across these studies , so subgroup analysis was carried out next. These results were presented in Figure Data from 107 studies were analyzed in a random-effect model to compare circulating adiponectin levels in people with different cancers and controls. Results showed that circulating adiponectin levels in cancer cases were significantly lower than in the controls with a pooled SMD of −0.334 μg/ml , which is consistent with the results derived from total adiponectin levels. Statistically significant amount of heterogeneity was observed across these studies . These results were presented in Figure HMW-adiponectin is the dominant form of adiponectin in plasma and correlates with cardiovascular disease, insulin resistance, and obesity , 114. BuStratified subgroup analysis was performed to evaluate the potential sources of heterogeneity including ethnicity, cancer type, study design, blood sample, assay method, study size, study quality and mean age of cancer patients Table . Lower lAdditionally, adiponectin was significantly lower in patients who used serum as test samples , and in 37 studies who used plasma as testing samples, 26 studies showed the inverse relation of adiponectin to cancer risk. Assay method (radioimmunoassay or enzyme-linked immunosorbent assay) did not affect the results that circulating adiponectin was lower in cancer patients with pooled SMD of −0.316 and −0.266. Study size (more or less than 100 patients) did not change the result of estimated SMD either . Besides, no matter the mean age of cancer patients is older or younger than 60 years, decreased adiponectin levels were still exist in cancer patients .Next we performed meta-regression to evaluate the effect of the above factors on the estimate of SMD. In meta-regression, none of the examined factors, such as ethnicity, cancer type, study design, blood sample, assay method, study size, study quality and mean age of cancer patients was proved to be significant contributing factors.Sensitivity analysis was performed by excluding one study at a time and calculating the pooled SMDs for the remaining studies. It was found that the combined SMDs were similar to one another and statistically significant. None of the studies influence the pooled results substantially in this analysis Table .P = 0.123).Publication bias was assessed by funnel plot and Egger's regression test. Funnel plot shapes demonstrated a marginally asymmetrical distribution Figure , accordiBy integrating 107 studies, our meta-analysis revealed that lower circulating adiponectin levels were associated with higher risk of cancers. Despite the existence of heterogeneity, the disparity of adiponectin levels between malignant individuals and controls reveals the potential ability of adiponectin to serve as a biomarker for early detection of cancers.Aberrant adiponectin secretion is associated with tumor progression, metastasis and overall prognosis. Two previous meta-analysis indicated that lower adiponectin levels were associated with higher risk of breast cancer, colorectal cancer and colorectal adenoma , 116. ByCirculating adiponectin levels are affected by various factors, including inflammatory, dietary, hormonal, genetic, and medicine. One of possible explanations for decreased adiponectin levels in malignancies is the sustained inflammatory status of cancer patients leads to the increased proinflammatory cytokines such as TNF-α and IL-6, which are all reported to suppress adiponectin transcription and translation in adipocyte cell line , 119. BeHowever, elevated adiponectin levels are also reported in hepatocellular carcinoma. Since adiponectin is mainly degraded in the liver and adiponectin levels are elevated in advanced disease including cirrhosis and virus-related cancer , 122. OnThe peripheral functions of adiponectin are mainly mediated through AdipoR1 and AdipoR2. The expression levels of AdipoRs vary between malignant tissues and their peritumoral normal counterparts. The upregulation of AdipoR1 and AdipoR2 are reported in gastric carcinoma , whereasDespite the inverse correlation between adiponectin and various cancers, the underlying mechanisms of adiponectin in potential cancer suppression are still need to elucidate. Adiponectin decreases low density lipoprotein (LDL) receptor expression in breast cancer cells through promoting autophagic flux and inhibits LDL-cholesterol-induced tumor cell proliferation . AdiponeOur study has some limitations that need to be addressed when interpreting the results. The significant heterogeneity was observed among the studies thus the conclusion should be more conservative. Although stratified analysis was conducted, none of the factors including ethnicity, cancer type, study design, blood sample, assay method, study size, study quality, and mean age of cancer patients were confirmed to contributing factors. Some possible reasons may partially explain this heterogeneity. Adiponectin levels are changed along with the tumor development. The tumor type, size, histological grade, and lymph node metastasis are the possible contributors caused heterogeneity. It is difficult for us to acquire the detailed information from the included studies. Besides, the subjects were from different regions and the lifestyle combined with diet was varied, which might influence the level of adiponectin. Since adiponectin is mainly secreted from adipose tissue, variables such as age, hormone receptor expression, menopausal status and BMI could contribute to the secretion and those factors were not fully deliberated for the complexity of tumor environment.In summary, the present study shows significant difference in circulating adiponectin levels between patients with malignancies and controls. Low circulating adiponectin level is associated with increased cancer risk, which suggests that adiponectin may serve as a potential biomarker for early detection of cancers considering its abundance in blood. Thorough understanding the roles of adiponectin and its receptors in the progression of cancers is helpful to cancer screening and promote individualized treatment.Based on the standard guidelines, a systematic search of English literature from Cochrane library, Wiley online library, PubMed was conducted to retrieve eligible studies until August 8, 2015. Searching terms included Medical Subject Heading (Mesh) and free text words “adiponectin”, “ADPN”, “Acrp 30”, “AdipoQ”, “GBP 28” or “apM1” in combination with “neoplasm”, “cancer”, “carcinoma”, “malignancy” or “tumor”. Furthermore, we manually searched references of relevant studies to add potential research to this meta-analysis.Studies were included if they met the following criteria: (i) full text case-control studies published in peer-reviewed journals evaluating the relationship between circulating adiponectin concentration and carcinogenesis; (ii) all cases were diagnosed as cancer by pathological biopsy or other medical methods with blood sample obtained before any therapies and all the controls were people without any cancers. (iii) circulating adiponectin level and standard deviation (SD) of it were provided or there were enough information to estimate them. Reviews, letters or animal experiments were excluded and articles without key information to carry on further analysis were also beyond consideration. Meanwhile, if replicated patient cohort was published in different studies, only the most recent or complete one was chosen. Since all the studies included were acquired from literature, ethics committee approval was not needed.Based on the checklist of MOOSE  , two revTwo reviewers (Tai W and Peng Y) independently assessed the quality of each included study according to the Newcastle-Ottawa Quality Assessment Scale (NOS)  ranges fP < 0.05 was considered statistically significant. Higgins I-squared statistic was applied to offer evidence of heterogeneity with I2 > 50% suggesting significant heterogeneity. The pooled SMD and 95% CI was calculated using a fixed-effects model if the heterogeneity was not significant, otherwise a random-effect model was employed and subgroup analyses and meta-regression were adopted to detect the potential cause of heterogeneity.We acquired the mean ± SD of circulating adiponectin levels from cases and controls through three ways. The most accurate method was extracted them from the original research directly. However, a few studies presented the results as median values or standard error. In that case, we regarded median value as mean value considering the large sample size and calculated the SD value by using standard error and population number. If necessary, we contacted the author for detailed information. Standard mean differences (SMDs) and the corresponding 95% confidence intervals (CIs) of circulating adiponectin were calculated for all the eligible studies. Cochran's Q-test was performed to test the heterogeneity of included studies and P values were two-sided.Sensitivity analysis was executed to detect the robustness of the results. Publication bias was evaluated by use of funnel plot and Egger's linear regression test. The Stata 13.0 software  was used to perform all the statistical analysis. All


Germline HOXB13 G84E mutation (rs138213197) has been described associated with prostate cancer (PCa) susceptibility but results of different studies are inconsistent. We conducted this meta-analysis to evaluate the specific role of this mutation. Relevant available studies were identified by searching the databases Pubmed, Embase and Web of Science. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to measure the strength of the association. Subgroup analysis were performed to evaluate the specific role of rs138213197 in disease aggressiveness, diagnostic age and family history. Furthermore, trial sequential analysis (TSA) was conducted for the first time to estimate whether the evidence of the results is sufficient. Our results indicated that significant increased PCa susceptibility was associated with rs138213197 compared with non-carriers . Besides, in subgroup analysis, HOXB13 G84E variant was obviously associated with early onset , affected relatives  and highly aggressive disease . By TSA, the findings in the current study were based on sufficient evidence. Therefore, our results indicated that the G84E mutation in HOXB13 gene might increase susceptibility to PCa. Prostate cancer (PCa), as the most common cancer among men, is the second leading cause of tumor death in men of USA , 2. VariGermline Homeobox B13 (HOXB13) encodes a transcription factor which belongs to the highly conserved homeobox gene family. Many studies have demonstrated the role of HOXB13 in embryonic development and tumor suppression , 12. In In most studies, rs138213197 has been revealed to be a significant risk factor associated with PCa incidence , 27–29. All remaining articles were checked to prevent overlapping datasets Figure . A totalSensitivity analysis was utilized to detect the influence of each study on the pooled OR by repeating the meta-analysis while omitting one single study each time. Our sensitivity analysis did not indicate alterations in the results if any individual study was excluded Figure .Overall, the pooled OR of the enrolled 16 studies was 3.38 (95% CI: 2.45–4.66) in a random-effect model, which indicated a strong association between HOXB13 G84E mutation and PCa susceptibility Figure . FurtherFor subgroup analyses of the age, family history and aggressive, all results showed a significant association between HOXB13 G84E mutation and PCa susceptibility    was conducted for a more comprehensive assessment in our current meta-analysis. Although the number of the cases and controls have not achieved the required information size, the cumulative Z-curve has crossed the monitoring boundaries already Figure , demonstMutations in key tumor suppressor genes, oncogenes, or mismatch repair genes might play a vital role in tumor occurrence and development . RecentlMeta-analyses have a greater power than a single study by combining all eligible studies. In the current study, there existed several advantages: (1) The sample size is larger than any single study, making our results more reliable and precise. (2) The Begg's test and sensitivity analysis showed no publication bias and low-quality study. (3) Subgroup analyses were conducted to explore the association between HOXB13 G84E allele and PCa susceptibility in different ethnicities, source of controls, genotyping methods, tumor stages, blood PSA levels, and Gleason scores. (4) Our results were based on sufficient evidence, which was proved by TSA for the first time.In the present meta-analysis, the results showed the mutation is associated with a 3.38-fold increased risk of PCa in total. Moreover, subgroup analyses in different ethnicities, source of controls and genotyping methods were conducted. Results of these subgroup analyses indicated significant association between rs138213197 and PCa risk. Noticeably, subgroup analyses were also performed according to disease aggressive and all the results indicated significant association, patients with more aggressive disease are more likely to carry the mutation. In conclusion, the present results indicated that the G84E mutation in HOXB13 gene might increase susceptibility to PCa.TSA is an approach that combines an a required information size with the adaptation of monitoring boundaries to evaluate the accumulating data . In our However, some limitations should also be emphasized. (1) Most populations included in this meta-analysis were Caucasian ethnicity, and more populations from other ethnicities will be required in future research. (2) The number of included studies in some subgroups was relatively small, with limited statistical power to investigate the real association. (3) Adjusted estimates could not be performed in our analysis without enough data for the adjustment by other PCa covariates such as age, life-style and so on. (4) Though TSA was conducted for the first time in this meta-analysis to assess the risk of random error, the shortage of TSA itself could not be avoided. More studies by standardized unbiased methods are required, which can offer more detailed individual data of high quality. (5) In our meta-analysis, the source of Gleason score and T-stage were different in various included studies, which might induce potential inaccuracy.This meta-analysis was strictly performed according to the preferred reporting items of the systematic reviews and meta-analysis (PRISMA) statement .Online databases including Pubmed, Embase and Web of Science were searched to identify relevant literature published until 20 April 2016. The combination of the following search items were utilized: “Germline Homeobox B13” or “HOXB13”, “polymorphism” or “rs138213197” or “G84E” or “p.Gly84Glu”, and “Prostate”. Additional eligible studies were searched manually from reference of original studies or reviews. Following criteria were utilized to select the published literature: (1) English publications; (2) case-control studies; (3) studies concerning the association between HOXB13 G48E variant and PCa risk.To maintain the quality of the meta-analysis, studies were excluded when: 1) no clear definitions of the study design, population, country of origin or the outcome assessment; (2) no sufficient data for estimating an odds ratio (OR) with 95% confidence interval (CI); (3) without control series; (4) duplicates of previous publication. Sensitivity analyses were also performed to avoid biases in the results due to certain low-quality studies  independently to determine whether an individual study was eligible for inclusion. The information were recorded in a standardized form and the extracted elements included: first author's name, year of publication, control source [population based (PB) or hospital based (HB)], ethnicity, genotyping method, the number of HOXB13 G84E mutation carriers and non-carriers respectively. All of the aforementioned data are comprehensively detailed in Table 2 statistic and Cochran Q test were applied to assess the heterogeneity [p < 0.05 or I2 > 50%); otherwise, the fixed-effect model was applied. An OR value greater than 1 indicated HOXB13 G84E variant was associated with higher PCa susceptibility and was considered statistically significant if the 95% CI did not include 1. After that, subgroup analysis was further carried out by diagnostic age, family history and disease aggressiveness. Early onset was defined as age ≤ 55 years in five papers [p < 0.05 was considered statistically significant. The meta-analyses were carried out with Stata12 .Sensitivity analysis was conducted firstly by calculating the results again by omitting one single study each time. The pooled odds ratios (ORs) with 95% confidence intervals (CIs) were utilized to evaluate the strength of association between rs138213197 and PCa susceptibility. Random-effect pooled ORs and 95% CIs were calculated through the DerSimonian–Laird method and the fixed-effect model based on the Mantel–Haenszel method according to heterogeneity among the pooled studies . Higginsogeneity . A randoe papers , 19–21, e papers , 23, 28 e papers , 27. Adde papers . p < 0.0Random error can mislead results in meta-analyses. The risk of random error may increase remarkably because of multiple looks on accumulating evidence when new trials emerge , 39. To In our study, 5% risk of a type I error and 10% risk of a type II error (a power of 90%) was set in TSA. Furthermore, 10% relative risk increase was predetermined according to the required information size and 90% confidence intervals was provided. The trial sequential analysis software  was utilized in this meta-analysis.Our results indicated that the G84E mutation in HOXB13 gene might increase susceptibility to PCa. Moreover, the effect size of HOXB13 G84E in PCa is more obvious in patients with relatives affected, more aggressive PCa or early-onset PCa.


This systematic review and meta-analysis is to evaluate the risk of development of concomitant strabismus due to refractive errors. Eligible studies published from 1946 to April 1, 2016 were identified from MEDLINE and EMBASE that evaluated any kinds of refractive errors  as an independent factor for concomitant exotropia and concomitant esotropia. Totally 5065 published records were retrieved for screening, 157 of them eligible for detailed evaluation. Finally 7 population-based studies involving 23,541 study subjects met our criteria for meta-analysis. The combined OR showed that myopia was a risk factor for exotropia . We found hyperopia had a dose-related effect for esotropia . Sensitivity analysis indicated our results were robust. Results of this study confirmed myopia as a risk for concomitant exotropia and identified a dose-related effect for hyperopia as a risk of concomitant esotropia. Strabismus, a misalignment of both eyes, is a common ocular disorder in childhood populations. The prevalence estimates for concomitant strabismus ranged from 2.3% to 6.0% in children13548The pathogenesis of different types of concomitant strabismus has not been well established. Many researchers have studied the association between accommodative esotropia and hyperopia111114http://www.nlm.nih.gov/bsd/licensee/medpmmenu.html) and EMBASE (available in the public domain at http:www.embase.com/info/helpfiles/) database from their starting date to April 1, 2016. The Boolean logic was adopted in the search strategy. The following keywords were used as free words and also as MeSH terms: “strabismus”, “esotropia”, “exotropia”, “refractive errors”, “myopia”, “hyperopia” and “anisometropia”. Detailed search strategies were given in Literature search was performed via the Ovid platform in MEDLINE (available in the public domain at The inclusion criteria for eligible studies were as follows: (1) population-based cross-sectional study; (2) the status of strabismus (heterotropia) was diagnosed by cover and uncover test; (3) refractive error of both eyes was documented; (4) the numbers or frequencies of strabismus and non-strabismus in each type of refractive error were reported, or odds ratio (OR) and 95% confidence interval (95% CI) of different refractive errors for strabismus were estimated. We excluded animal studies, case reports, reviews, abstracts, conference proceedings, editorials, non-English articles and reports with incomplete data.All retrieved records were screened and reviewed by two independent reviewers (SMT and RYTC). Uncertainties were resolved by consensus with a third reviewer (JCSY). Data collected from each study included (1) study information including year of publication, country of study, age range of participants, definition of different refractive errors , sample sizes; (2) numbers of strabismus in subjects with and without refractive errors, reported unadjusted and adjusted ORs and 95% CIs (or standard errors), and adjusted co-variables; and (3) numbers of esotropia and exotropia in subjects with and without refractive errors, if provided.All included studies were population-based cross-sectional studies. Therefore, the quality of studies was assessed via the modified Estabrooks’ Quality Assessment and Validity Tool2 statistic quantifying the proportion of total variation attributable to between-study heterogeneity were calculated2 above 50% indicated high heterogeneity. If significant heterogeneity was detected, result from the random-effect model was adoptedWe evaluated the association between concomitant strabismus and refractive errors by synthesizing the outcomes using meta-analysis. Among the eligible studies, four studies have classified concomitant strabismus into esotropia and exotropia111217From 1946 to April 1 of 2016, a total of 6962 publications were identified from the EMBASE and MEDLINE databases. After detailed screening and evaluation, 157 reports were eligible for detailed evaluation. Among them, we found 7 articles11112171823111217118111217121712172 = 65%, P < 0.0001; 11122 = 69%, P = 0.0001; 2 = 43%, P = 0.1; 2 = 0%, P = 0.05) (Six out of seven studies111121823 = 0.05) . Only Zh2 = 95%, P = 0.002; 11122 = 81%, P < 0.0001; 2 = 97%, P = 0.32; Six out of seven studies1111218232 = 76%, P < 0.0001; 11122 = 0%, P < 0.0001; 2 = 3%, P < 0.0001; Six out of seven studies1111218232 = 92%, P < 0.0001; 1112172 = 72%, P < 0.0001; 2 = 83%, P < 0.0001; Six out of seven studies1111218232 = 31%, P < 0.0001)All studies were of high quality indicating low risk of bias when being included in this meta-analysis . Egger’sThis systematic review and meta-analysis on 7 large-scale population-based studies involved 23,541 children to determine the association of refractive errors and concomitant strabismus. Based on the synthesized unadjusted ORs, all types of refractive errors  were confirmed as risk factors for concomitant strabismus.Among the 7 studies, three have studied the association of myopia with exotropia and esotropia separately. Our results demonstrated no significant association between myopia and esotropia. On the other hand, children with myopia had 5.23-fold increase in risk to develop exotropia than those without significant ametropia. This may explain the high prevalence of exotropia in Asia, in which the prevalence of myopia is much higher111528Our results revealed a very strong association between hyperopia and esotropia in a dose-related effect manner. We have therefore identified increasing risk of developing concomitant esotropia with the severity of hyperopia. Based on the meta-analysis of two papers12When pooling up the unadjusted ORs of three articles, which have separated into esotropia and exotropia, astigmatism was also found to be a risk for both esotropia and exotropia. However, the results need to be interpreted with cautions, because the confounding effect of spherical myopia and hyperopia has not been adjusted in the analysis. In the seven studies for this meta-analysis, only two reports have provided the adjusted ORs for the effects of astigmatism to exotropia. No association was found between astigmatism and exotropia based on the pooled adjusted ORs. Therefore, further studies were needed to confirm the association.Anisometropia was found to be associated with strabismus and both esotropia and exotropia based on pooled unadjusted ORs. This can be attributed to the much reduced binocularity in children with anisometropiaIn this meta-analysis, all included studies were population-based studies. The results obtained are of epidemiological relevance. Furthermore, we used risk of bias assessment tools for observational study recommended by MOOSE guidelines and Cochrane handbook for Systematic Reviews, which showed all included studies were of good quality. Sensitivity analysis has been conducted to further confirm our findings and no significant publication bias has been found. However, all the 7 reports were cross-sectional studies. Further longitudinal studies are warranted to establish a causal relationship between the two conditions. Moreover, we found the definition of refractive errors varied among studies, which could be a major source of heterogeneities. In our meta-analysis, we accounted for heterogeneities by using random-effect model and sensitivity analysis.In summary, we confirmed that myopia increased the risk of exotropia. Hyperopia was associated with increased risk of esotropia in a dose-dependent manner. We also reported a suggestive association of astigmatism and anisometropia with concomitant strabismus, which should be further confirmed in follow-up studies. Refractive errors are extremely common especially in Asian populations. Strabismus is a difficult ophthalmic disorder that disrupts vision and depreciates quality of life. Investigations are warranted to understand the pathophysiology for the associations between refractive errors and concomitant strabismus.How to cite this article: Tang, S. M. et al. Refractive Errors and Concomitant Strabismus: A Systematic Review and Meta-analysis. Sci. Rep.6, 35177; doi: 10.1038/srep35177 (2016).


Systemic lupus erythematosus (SLE) is a chronic systemic inflammatory autoimmune disease, the etiology of which remains only partially characterized. Strong evidence implicates chronic infections in the development and chronicity of autoimmune conditions. Recently, an association has been demonstrated between periodontitis and rheumatoid arthritis. Such observations have led to the investigation of the possible role of periodontitis and oral dysbiosis in other systemic inflammatory conditions, including SLE. The aim of this study was to examine whether there is an association between SLE and periodontitis.via OVID, EMBASE via OVID, and PsycINFO via OVID databases were searched to identify eligible studies, screened by two independent authors and verified by a third. Studies comparing presence of periodontitis in SLE cases to controls without SLE were included. Data were extracted using a predefined table and papers were appraised using Down’s and Black tool. Mantel–Haenszel meta-analysis was performed using RevMan.MEDLINE p = 0.0004). No statistical difference was found in individual measures of periodontitis, such as probing depth or clinical attachment loss, between SLE cases and controls.Eight case–control studies were included, with 487 SLE cases and a total of 1,383 participants. On meta-analysis of four studies, risk of periodontitis in SLE cases compared to controls was significantly greater with a risk ratio of 1.76 (95% CI 1.29–2.41, Our study found a statistically significant increased risk of periodontitis in patients with SLE compared to controls. This finding suggests a possible association between these two conditions. Larger longitudinal studies are needed to confirm this possible association. Systemic lupus erythematosus (SLE) is a systemic, chronic inflammatory condition with diverse clinical manifestations, primarily affecting the joints, internal organs, and the skin .The etiology of SLE is incompletely understood, but it is thought to occur in genetically primed individuals in whom the inflammatory response is triggered by an environmental stimulus. Immunosuppressant medications are the mainstay of treatment, but these are limited both in efficacy and by multiple side effects which lead to significant morbidity and mortality. Oral manifestations of SLE are common and typically take the form of painless oral ulcers that are frequently present during disease flares and are included in current SLE classification criteria .Porphyromonas gingivalis, Tannerella forsythia, and Treponema denticola, have a key role in the development of periodontitis . A recent meta-analysis including 153,492 participants showed a significant association between periodontitis and rheumatoid arthritis . Two spengivalis , 9 and Acomitans . ClinicaThere is increasing interest regarding the possible role of oral dysbiosis in the etiology of other autoimmune inflammatory conditions, including SLE. Further understanding of any potential association between periodontitis and SLE would expand current knowledge of the etiology of SLE and may lead to novel management strategies.In this review, we hypothesized that there may be an association between SLE and periodontitis. To evaluate this hypothesis, we conducted a systematic review and meta-analysis of relevant publications.http://www.crd.york.ac.uk/PROSPERO/). We used the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) 2009 checklist to report the review (The protocol for the review was registered with PROSPERO (Registration number: CRD42016053490) an international register of systematic reviews , probing pocket depth, bleeding on probing (BOP), and clinical attachment loss (CAL). In addition, Google Scholar was searched using the following term “Periodontitis and systemic lupus erythematosus.” The searches were re-run just before the final analyses on the 03/08/2017.MEDLINE The Titles and/or abstracts were screened by two independent authors (Zoe Rutter-locher and Toby O. Smith) to identify potential eligible studies. Final selection of studies was performed by two independent authors (Zoe Rutter-locher and Toby O. Smith) and verified by a third author (Nidhi Sofat) by reviewing the full text based on inclusion criteria above. Any disagreements were resolved by discussion.A standardized, pre-piloted form was used to extract data from the included studies. Extracted information included: (1) study design including, publication journal and date, inclusion and exclusion criteria, diagnostic criteria of SLE, definition of periodontitis, (2) study participant demographics including % females, mean age, years of SLE disease, therapies, measure of severity of SLE, and (3) periodontal measures including prevalence of periodontitis, oral plaque index (PI), probing depth (PD), clinical attachment loss (CAL), and bleeding on probing (BOP). Two authors (Zoe Rutter-locher and Toby O. Smith) extracted the data independently, discrepancies were identified and resolved through discussion . Authors were contacted by email to obtain missing information and included when received.Risk of bias and quality assessment was reviewed using Downs and Black tool for non-randomized control trials . This 27As there was homogeneity in participants, study design and outcome measure on visual assessment of the data extraction table, a meta-analysis was performed. Primary outcome was to calculate relative risk of periodontitis in participants with SLE compared to participants without SLE. Secondary outcomes were to calculate relative risk or mean difference for measures of periodontal disease. These measures of periodontal disease included PD, oral PI, BOP, bleeding gingival index, and CAL.I-squared) was ≤50% and Chi-squared equates p = 0.10 and a random-effect meta-analysis when I-squared was >50% and Chi-squared equates to p < 0.10. Risk ratio with 95% confidence intervals was calculated for the prevalence rates of periodontitis and the mean difference was calculated for continuous variables. Risk of bias was identified in Down’s and Black tool. Publication bias was not performed as it is convention to only present a funnel plot for 10 or more data points in a meta-analysis (Median and interquartile range were converted to mean (SD) to allow comparison of studies, under the assumption of normal distribution. A fixed effect meta-analysis was performed when the inconsistency value .As shown in Figure The study characteristics are shown in Table The total number of study participants was 1,383 which included 487 cases of SLE. All studies defined cases as fulfilling the ACR 1982/1997 revised classification criteria for diagnosis of SLE . Cases wN = 8, 100%). The participants were representative of the population  and the outcome measures were reliable and clearly reported  with estimates of random variability . Particular limitations were that assessors were not blinded  and sufficiently powered cohort size was present in only half of studies .Quality assessment was hindered by the limited information available for some studies, especially those that came from conference proceedings. As shown in Table Table As shown in Table p = 0.0004, Figure p = 0.02) and when the study by Wang et al. was excluded from meta-analysis . Interestingly two studies which differentiated between mild and severe periodontitis found no difference in the prevalence of severe periodontitis between cases and controls, whereas one did (The prevalence rates of periodontitis were similar in these four studies. The exceptions were the lower prevalence rate of periodontitis in the controls of the study by Wang et al. and Zhang et al. , 22. On All studies provided data on measures of periodontal disease. Six of the eight studies reported oral PI , 22, 23.p = 0.62, Table p = 0.18, Table Number of residual teeth was very similar in all studies. Overall, the mean residual teeth count was 23.68 for SLE and 23.65 for controls. On meta-analysis there was no significant difference in mean of PI (0.03, 95% CI −0.09–0.16, p = 0.06, Table p = 0.08, Table p = 0.66, Table p = 0.13, Table We also performed meta-analysis on the gold standard measures of periodontitis. Risk of PD ≥ 5 mm was greater in SLE patients compared to controls but this was not statistically significant with risk ratio of 1.34 (95% CI 0.99–1.82, Our report is the first systematic review to examine the association between periodontitis and SLE. On meta-analysis we found a statistically significant overall increased risk of periodontitis in patients with SLE compared to controls, suggesting an association between these two conditions. However, there was no statistical difference in individual measures of periodontitis, such as PD or CAL, between SLE cases and controls.We found no significant difference in oral PI or BOP between SLE cases and controls. Plaque induced periodontitis is the most common form of periodontitis, and so a high oral PI would suggest a greater risk of developing periodontitis. BOP is a measure of gingivitis, the reversible gingival inflammation which precedes periodontitis. Therefore, although these measures do highlight potential risk of periodontitis, the absence of a significant difference between cases and controls does not preclude an association between SLE and periodontitis.There was also no difference in means of PD. PD calculates the depth of periodontal pockets and is a measure of current periodontal disease. The British Society of Periodontology delineates a healthy sulcus as <3.5 mm and a periodontal pocket as ≥3.5 mm . We propClinical attachment loss is the other gold standard measurement for periodontitis  and is rAggregatibacter actinomycetemcomitans, which has been identified as a potential trigger in RA (The finding that periodontitis is associated with SLE is in agreement with other studies examining this relationship. Case studies since the 1980s have suggested a link between SLE and gum disease. Rhodus and Johnson found 93.8% of SLE patients had periodontitis , while aer in RA , has alser in RA .Systemic lupus erythematosus is thought to occur when an environmental stimulus triggers inflammation in a genetically primed individual. Initial studies have highlighted possible mechanisms to explain the potential association between periodontitis and SLE. Genetic variants in the Fcy receptor have been implicated in susceptibility to both SLE and periodontitis in a small Japanese study . PeriodoA significant limitation to our meta-analysis is the lack of clear and recognized criteria to define periodontitis. Although all definitions of periodontitis used PD and CAL, they did vary in their thresholds, limiting the ability to directly compare outcomes. However, a minimum diagnostic threshold of CAL ≥ 2 mm and PD ≥ 3 mm has been suggested  and all Another significant limitation is the paucity of data available. Only eight studies could be included in the meta-analysis. The overall periodontitis risk ratio included only four studies, and one of these was from conference proceedings. Meta-analysis involving greater number of studies analyzing individual measures of periodontitis were not significant. This may suggest that the significant risk ratio in overall periodontitis cannot be extrapolated into larger study populations. However, previous similar meta-analysis in other conditions, which have included larger numbers of studies, have also found that overall periodontitis risk is significant, while individual markers are not .There are only a small number of studies to date investigating this association and so we were unable to test for publication bias. We included data from conference proceedings try to mitigate this but it will be important in the future to ascertain the risk of small sample size publication bias.There are a number of factors such as smoking, educational level, and immunosuppressant medications which increased risk of both SLE and periodontitis . VariatiFinally, all studies included were cross-sectional in nature and investigated an association at a given time point. We are, therefore, unable to make any conclusions regarding causality from this study. Further longitudinal studies are needed to delineate a temporal association and causality of periodontitis in the development of SLE.The results of this meta-analysis show a significant association between SLE and periodontitis. However, the meta-analysis was hampered by paucity of data and significant limitations. Therefore, these findings can only suggest a possible association and larger longitudinal studies are needed to confirm this association and investigate causality of periodontitis in SLE.ZR-l, TS, and NS conceived, analyzed, and drafted the manuscript. IG conducted literature searches and reviewed the manuscript. All authors approved the manuscript for publication.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


As there is conflicting evidence for the relationship between hepatitis B virus surface antigen (HBsAg) positivity and the prevalence of metabolic syndrome (MetS), we performed a meta-analysis to investigate whether HBsAg positivity affects the incidence of MetS.Observational studies on the relationship between HBsAg positivity and MetS were obtained from PubMed, Web of Science, and the Cochrane Library in April 2016. The pooled odds ratios (ORs) of MetS and its components  for subjects with or without HBsAg positivity were synthesized. The standardized mean difference of MetS components between HBsAg-positive participants and healthy controls was calculated. Heterogeneity was explored with subgroup analysis and sensitivity analysis. Publication bias was detected using Egger’s test and Begg’s test.Thirty studies were eligible for meta-analysis. The MetS OR for HBsAg-positive participants was significantly decreased compared with the controls . The negative effect of HBsAg positivity on elevated triglycerides  was strong, while that for increased fasting blood glucose was weak . The pooled ORs of central obesity , reduced high-density lipoprotein cholesterol , and elevated blood pressure  for HBsAg-positive participants were all not significantly different compared with the controls. No publication bias was detected.Serum HBsAg positivity is inversely associated with the prevalence of MetS. Among the five components of MetS, elevated triglycerides had the strongest inverse relationship with HBsAg positivity. Chronic hepatitis B virus (HBV) infection remains a globally challenging problem, as it can lead to chronic active hepatitis, liver cirrhosis, and hepatocellular carcinoma , 2. MetaThe liver plays an undeniably important role in lipid and glucose metabolism. MetS involves dyslipidemia and glucose abnormalities. Dyslipidemia is associated with the development of obesity and hypertension, which are also components of MetS. Additionally, nonalcoholic steatohepatitis is considered the hepatic manifestation of MetS , 6, and HBV and hepatitis C virus (HCV) are two common types of hepatitis virus that share some similarities. Chronic HCV infection contributes to MetS, as it induced insulin resistance in a genotype-dependent model . HoweverThis meta-analysis was performed according to a proposal for reporting meta-analysis of observational studies . We searStudies were deemed eligible if they met the following criteria: (1) investigated the association between HBsAg positivity and MetS . HBV infection was defined as HBsAg seropositivity; (2) used healthy subjects as the control group; (3) included >30 subjects with HBsAg positivity; otherwise, a study was excluded for low statistical power and poor reliability. Exclusion criteria were studies on co-infection, such as human immunodeficiency virus and HBV co-infection, liver cirrhosis, hepatocarcinoma, following antiviral therapy, pregnant or pediatric populations.Two authors (L.Y.Y. and Z.Y.) independently assessed the quality of eligible studies. The Newcastle-Ottawa Scale criteria  were recFor continuous variables, the mean and standard deviation (SD) of WC, body mass index (BMI), TG, HDL-C, FBG, systolic BP, and diastolic BP for HBsAg positive subjects and the controls were extracted. For categorical variables, the adjusted OR was extracted; otherwise, the crude data were extracted to calculate the OR. In addition, the datasheet included the publication year, region, study design, source of subjects, sample size, mean age, gender distribution, and diagnostic criteria of MetS.2 test. The degree of heterogeneity was classified to three levels  [P > 0.05 and I2 < 50%), the fixed effect model was used to calculate the pooled OR and 95% CI. Otherwise, the random effect model was used. To investigate the source of heterogeneity, subgroup analysis and sensitivity analysis was performed according to the factors related to quality assessment. Publication bias was assessed with Egger’s test [P < 0.05). Statistical analyses were conducted with Review Manager 5.3 (The Cochrane Collaboration) and STATA 11.0 .The standardized mean difference (SMD) of WC, BMI, TG, HDL-C, FBG, systolic BP, and diastolic BP between the HBsAg-positive group and controls was calculated. Then, the pooled SMD and associated 95% confidence intervals (CI) were obtained from a DerSimonian and Laird random effects model . More im2 ≥ 50%) . If no sr’s test  and BeggWe retrieved 2687 studies using the described search strategies. We excluded 2657 studies in accordance with our inclusion and exclusion criteria . Ultimatet al. [et al.  consideret al. , 27, 42 et al. . The Q7,2 = 72%, P < 0.01) . The pooled OR from nine studies [2 = 77%, P < 0.01). After excluding the confounder of HCV, the pooled OR was 0.70 . The pooled OR from data adjusted for confounders was 0.73 . This inverse association was also found in the male subgroup , but not in the female subgroup . Furthermore, the heterogeneity did not decrease through subgroup analysis, therefore the specific factor leading to heterogeneity was not found.Twelve studies , 32, 33  < 0.01) , indicat studies , 32, 33 2 = 50%, P = 0.03) . The heterogeneity also decreased in subgroups stratified by sex; the conclusion was identical to the total pooled OR , which was consistent with WC.WC and BMI are two common indices for assessing central obesity. Eleven studies , 33, 43  = 0.03) , indicatooled OR . Additio2 = 0%, P = 0.52) , indicating that the HBsAg-positive subjects had lower TG than the healthy controls. Although the OR and SMD were calculated from different studies, they revealed a consistent trend.Fourteen studies , 33–35 i = 0.52) , indicat2 = 85%, P < 0.01) . The dramatic decrease in heterogeneity was due to adjusting for confounding factors . However, the dramatic decrease in heterogeneity rendered the results more reliable, the inverse relationship was still weak.Thirteen studies , 33, 34 .01) see , indicat2 = 95%, P < 0.001) . The subgroup that included seven studies [2 = 0%, P = 0.63). Additionally, similar trends were found in the SMD of systolic BP and diastolic BP. In conclusion, HBsAg positivity was neither a risk factor nor a protective factor for increased BP, and the difference in BP between HBsAg-positive subjects and healthy controls was not significant.Eleven studies , 33, 42 001) see . After e studies , 31, 33 2 = 21%, P = 0.23) . Six studies [2 = 45%, P = 0.11). The SMD derived from 16 studies [2 = 99%, P < 0.0001). Overall, the effect of HBsAg positivity on glucose homeostasis appeared slight. However, further research is required to confirm this.Thirteen studies , 33, 34 .23) see , indicat.23) see . Seven s studies , 27, 28  studies –47 was 0P > 0.05, so there was no statistical difference between publication bias and zero, meaning no publication bias was present; Begg’s test derived the same conclusion. Taken together, this indicates that there was no publication bias in our meta-analysis.Publication bias was not detected by Egger’s test or Begg’s test . For EggIn this meta-analysis, HBsAg-positive individuals had lower prevalence of MetS. This negative association remained robust after adjustment for confounding factors . Meanwhile, a strong inverse relationship was demonstrated between HBsAg positivity and elevated TG (one component of MetS). There was a slight effect of HBsAg positivity on glucose homeostasis. The total OR of all eligible studies indicated no association between HBsAg positivity and reduced HDL-C, but OR controlled for the confounding factors revealed a slight inverse relationship. Additionally, it was confirmed that HBsAg positivity is not associated with central obesity and increased BP. Overall, we speculate that HBsAg positivity protects against the incidence of MetS mainly due to its negative effect on elevated TG. Naturally, further research is required to confirm this.There was a negative association between HBsAg positivity and the prevalence of MetS, and HBsAg positivity is closely related to HBV. HBV may prevent the occurrence of MetS instead of promoting it. That is, HBV may protect humans against MetS. HBV is considered a “metabolovirus”, as it adopts a regulatory system that is unique to the major hepatic metabolic genes that control hepatic glucose and lipid metabolism . HBV infet al. [A recent review  has alsoet al.  reportedet al. . Besideset al. , and we et al. , 34, 35 et al. , 46, 47 et al. , 45–47 wet al. . In shorThe inverse relationship between HBsAg positivity and increased FBG was statistically significant, but was weak in the clinic. The relationship between HBV and insulin resistance remains inconclusive and awaits further studies for clarification . Howeveret al. [To the best of our knowledge, this is the first meta-analysis to investigate the relationship between HBsAg positivity and MetS (including its components). Additionally, this meta-analysis was performed rigorously according to a proposal for reporting meta-analysis of observational studies . Althouget al.  also focThere are several limitations to the present meta-analysis. First, the majority of eligible studies were cross-sectional studies, which always demonstrate the least evidence among the three types of observational studies . Additionally, time is an important factor that should be considered, as HBsAg-positive individuals may have different outcomes. Unfortunately, it was difficult to assess the impact of time in this meta-analysis, which we attribute to the cross-sectional nature of the included studies. Second, because only HBsAg was tested and/or it was tested for only once in most of the eligible studies, various conditions related to HBsAg were not taken into account. An HBsAg-positive individual may be a healthy carrier, a patient with chronic active hepatitis, or a patient with liver cirrhosis. Although most studies focused on the general population and most HBsAg-positive subjects may be HBV carriers in this meta-analysis, further stratification of HBsAg status is still needed to assess the exact role of HBsAg in the development of MetS in the future. Third, both age and gender play an important role in the natural history of chronic HBV infection. Unfortunately, the studies included in the subgroup analysis based on these two factors were very limited; however, the negative association between HBsAg passivity and MetS remained robust after adjustment for confounding factors . Fourth, with respect to the definition of MetS, we were not concerned whether drug treatment was an alternate indicator. Finally, we were unsuccessful in obtaining supplemental information from several authors; however, no publication bias was detected.Our meta-analysis has several implications for future research. First, a prospective large-cohort study is needed to validate our conclusion. In this regard, the Newcastle–Ottawa Scale  describeIn conclusion, serum HBsAg positivity is inversely associated with MetS. Among the five components of MetS, elevated triglycerides had the strongest inverse relationship with HBsAg positivity.S1 Checklist(DOC)Click here for additional data file.S1 Fig(TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.S3 Fig(TIF)Click here for additional data file.S4 Fig(TIF)Click here for additional data file.S1 Table(DOC)Click here for additional data file.S1 Text(DOCX)Click here for additional data file.