Numerous quantitatively based studies measuring the association between gout and the risk of fractures remain inconclusive. In order to determine whether gout could increase the risk of fractures, a meta-analysis was performed systematically.Electronic databases, MEDLINE/PubMed, Embase, and Cochrane Library were systematically searched to identify studies evaluating the association of gout and the risk of fractures. No restrictions on language, publication date, or journal of publication were imposed. Meta-analysis was performed to pool the outcome estimates of interest such as fracture incidence, fracture risk, and fracture risk in different sites and at different time points in the follow-up period.Screening determined that seven studies involving a total of 684,964 participants  were deemed viable for inclusion in the meta-analysis. The results of the analysis showed that gout would not significantly have a relatively higher risk of any fracture . Subgroup analysis showed consistent results for sexuality  and several occurring sites . Additionally, the results demonstrated that urate-lowering drugs prescribed early during disease had neither adverse nor beneficial effect on the long-term risk of fractures .This meta-analysis confirmed that gout was not associated with an increased risk of fractures. Urate-lowering drugs prescribed early during the course of disease had neither adverse nor beneficial effect on the long-term risk of fractures.The online version of this article (10.1186/s13018-019-1317-4) contains supplementary material, which is available to authorized users. Gout affected 2.4% of adults in the UK  and moreNumerous quantitatively based studies debating the association between gout and the risk of fractures remain inconclusive. Tzeng et al.  reportedThe aim of this study was to assess precisely the risk of fractures among patients with gout and estimate the potential effect of urate-lowering therapy on the risk. Furthermore, several subgroup analyses based on the gender, fracture sites, and etc. were performed to reveal more influencing factors.The checklist of the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement was followed in the conduction of this investigation  were searched through an iterative process using a combination of keywords and mesh terms “gout,” “arthritis,” or “uric acid” AND “fracture,” “bone fracture,” or “broken bone.” No restrictions on language, date, or journal of publication were imposed. Additionally, reference lists of related literature  about gout patients and associated fracture risk were carefully screened to retrieve additional eligible studies not identified by electronic database searching.Studies were included in the qualitative analysis if they confirmed criteria listed as below: (1) participants, subjects with a primary diagnosis of gout; (2) control group, subjects with no gout; (3) incidence of fractures was assessed in the gout group and control group; and (4) randomized/quasi-randomized/cluster controlled clinical trials and retrospective/prospective cohort studies.Exclusion criteria were (1) study protocols, letters, correspondence, and conference addresses; (2) experimental studies or animal studies; or (3) trials without sufficient data to obtain endpoint outcomes of interest.Following information was collected from included articles: first author’s family name, year of publication, study design, inclusion criteria of participants, source of data, number of participants in the gout group and the control group, demographic and clinical characteristics of participants , follow-up, evaluation endpoint outcomes of interest , and the drugs of urate-lowering therapy. Risk ratios (RR) and its related 95% confidence intervals (CIs) were directly extracted from original articles if provided or calculated from individual patients’ data.Newcastle-Ottawa Scale (NOS)  was usedTwo researchers (LFX and FX) repeatedly performed the processes of study screening and selection, data extraction, quality assessment of included studies, and data analyses independently. Any contradictoriness was resolved by discussion or an arbitrator (DJL) when consensus cannot be reached. Informed consent and ethics approvals were not required because all relevant data were extracted from published articles.I2 . Sensitivity analysis (conducted by omitting studies one by one) and publication bias (performed using Egger test) were implemented to evaluate the stability of results. A two-tailed p value < 0.05 indicated statistically significant.Meta-analysis was implemented to conduct the quantitative analysis and produce forest plots using Stata 12.0 Version . Evaluation parameters of interest involved the pooled RR as well as related 95% CIs was used to assess fracture risk between these two groups using random-effect models. Subgroup analyses were performed to explore the source of heterogeneity that was assessed using the A total of 2645 records were screened in the electronic databases searching process with another eight additional studies retrieved from screening reference lists of related articles. After 326 studies excluded by duplicates, 2291 excluded by reading titles and abstracts, and 32 studies were downloaded and carefully checked by reading full texts. After careful reading, five articles , 23–25 iAll included articles were published during 2016–2018 in English. Sample sizes of these studies ranged from 2674 to 292,808. With regard to the data source, all seven studies –26, 28 iAs for the risk of bias of included articles, six studies , 26, 28 The combined results as generated from five studies , 23–25 iThe combined results as generated from four , 24, 25 The pooled results as generated from three , 17, 23 The combined results as generated from four studies , 26, 28 This meta-analysis aimed to address the fracture risk of patients with gout and demonstrated that gout was not associated with an increased risk of fractures for any site, such as the humerus, wrist, vertebral, and hip fracture. In addition, urate-lowering drugs prescribed early during the course of disease had neither adverse nor beneficial effect on the long-term risk of fractures.As a typical feature of gout, high UA levels were demonstrated an inconsistent relationship with BMD and fracture, because the diverse properties of UA depend on the specific conditions. The protective effects of UA on BMD are mainly because of the antioxidant function or via vitamin D/parathyroid hormone pathway –31. The Hyperuricemia is a necessary but not sufficient predisposing factor for gout , which iOur research based on the analysis of seven studies including 151,002 gout patients and 533,962 participants in the control group revealed that there was no significant association between gout and fracture risk. Three of these articles demonstrated different conclusions from ours. Although Tzeng et al.  found thWe found that gout-related medication had a neutral effect on the risk of fractures, although previous literature on this topic was conflicting. Dennison et al.  reportedThe strength of this study compared with previous investigations lies in the large quantity of included data, reliable statistical methods, and comprehensive evaluation indexes. This meta-analysis included eight datasets by exhaustive database searching and reference screening, among which there were seven datasets from national or regional register databases. The large participant base made the results of this study well represented. Moreover, outcomes of interests were compared using a pairwise meta-analysis method and analyzed from multi-dimension to fully elucidate the association between gout and the risk of fractures and discuss the underlying mechanism. Findings of this meta-analysis provide the currently most comprehensive evidence about this issue, which will help gout patients and their clinicians to determine whether or not to make an appropriate therapeutic regimen to propose the fracture. Hence, the conclusions based on these articles were not fairly reliable.Nevertheless, this investigation was not without limitations. First, we only demonstrated an overall effect of gout on fracture but were not able to analyze the influence of some clinical and demographical factors . Second, there were only two prospective studies, and no randomized controlled trials were included. However, given the properties of gout and source of data, large scale RCT is currently inaccessible. Third, several studies did not present all characteristics of participants, and these missing data affect and weaken the strength of conclusions.This meta-analysis confirmed that gout was not associated with an increased risk of fractures. Urate-lowering drugs prescribed early during the course of disease had neither adverse nor beneficial effect on the long-term risk of fractures.Additional file 1:PRISMA checklist. The checklist of the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA). (DOC 64 kb)Additional file 2:Supplementary figures 1–3 and table 1. (DOC 2252 kb)


Preeclampsia is an idiopathic multisystem disorder with partial genetic and immunological etiology. Several studies investigated the association between various single-nucleotide polymorphisms (SNPs) in Fas and Fas ligand (FasL) genes and the risk of preeclampsia. However, they achieved inconsistent results. Therefore, we conducted a meta-analysis by systematically searching the Cochrane Library, PubMed and Embase databases and assessed this association by calculating pooled odds ratios with 95% confidence interval to reach a more trustworthy conclusion. Subgroup analyses by genotype methods and source of controls (SOC) were also conducted. Seven citations containing nine studies were included for four SNPs  in this meta-analysis. Our data suggested the G allele and genotype GG of the Fas -670 A/G polymorphism, GG genotype of the FasL 124A/G polymorphism, and TT genotype of the FasL -844C/T polymorphism increased the risk of preeclampsia. Stratification analyses by genotype methods and SOC also indicated that Fas -670 A/G polymorphism was related to increased risk for preeclampsia. In conclusion, Fas and FasL gene polymorphisms play important roles in the development of preeclampsia. Further well-designed studies in other races are needed to confirm the findings of this meta-analysis. Preeclampsia is an idiopathic multisystem disorder with partial genetic and immunological etiology . PreeclaRecently, several studies  reportedWe searched the Cochrane Library, PubMed and Embase databases to identify studies through August 30, 2018. The following key words were used: ‘Fas’ or ‘TNFRSF6/’ or ‘CD95’ or ‘APO-1’, ‘Fas Ligand’ or ‘FasL’ or ‘NFSF6’ or ‘CD95L’, ‘SNP’ or ‘polymorphism’ and ‘preeclampsia’ or ‘PE’. No restrictions were placed on the search. Additional initially omitted studies (such as reference lists of identified studies) were identified by hand screening.P value, and (4) case–control studies. Exclusion criteria were as follows: (1) incomplete data; (2) review or case report; (3) duplicate or overlapped publication. All questionable publications were discussed with consensus. Two reviewers independently screened the titles and abstracts.The identified studies conformed to the following criteria: (1) studies that evaluated the association between preeclampsia risk and Fas, FasL gene polymorphisms, (2) studies on human beings, (3) studies provided sufficient data to calculate the pooled odds ratios (ORs) and 95% confidence interval (CIs), and 2 test (http://ihg.gsf.de/cgi-bin/hw/hwa1.pl). The NOS criteria were scored according to three aspects: (1) subject selection: 0–4, (2) comparability of subject: 0–2 and (3) exposure: 0–3. The total NOS scores ranged from 0 (lowest) to 9 (highest). All disagreements were discussed and resolved with consensus.Related information was carefully extracted from included studies. The extracted information from all eligible studies including: author name, publication year, nationality, age, sample size, ethnicity, genotype methods, source of controls, and genotype numbers of cases and controls. Two reviewers independently performed the extraction of data and assessed the study quality based on the Newcastle–Ottawa Scale scores (NOS) . Hardy–WP<0.05 was considered statistically significant. Pooled ORs were calculated for all five gene models . If a Q-test indicated I2 < 50% or P>0.1 indicated heterogeneity across studies, a fixed-effect model was used. Otherwise, the random-effects model was used [P<0.05 was considered to indicate statistically significant.Stata 12.0 software  was used to perform all statistical analyses. We assessed the strength of associations between Fas, FasL genes polymorphisms and preeclampsia risk by ORs and 95%CIs. Stratification analyses were carried out by source of controls (SOC) and genotype methods. was used . We perfwas used ; P<0.05 We yielded a total of 108 citations after incipient search. Sixteen citations were selected for further full-text review. Nine citations were excluded due to the following reasons: two citations did not provide detailed genotyping data; four studied other diseases, and three was not case–control study. Eventually, we identified seven eligible citations  .P=0.014) for the 124A/G polymorphism increased the risk of preeclampsia. We also detected a significant association between FasL gene -844C/T polymorphism with increased risk for preeclampsia .Results of pooled analysis on the association between FasL gene 124A/G polymorphism and preeclampsia risk were shown in The Fas gene -1377 G/A polymorphism was investigated only in one study , which rIn this meta-analysis, our data showed that the Fas -670 A/G polymorphism, FasL 124A/G polymorphism, and FasL -844C/T polymorphism increased the risk of preeclampsia among Caucasians. Stratification analyses of genotype methods and SOC also yielded similar increased risks for preeclampsia.Some studies provided evidence that disturbances in apoptosis were associated with increased risk for preeclampsia ,19. Fas Some limitations in this meta-analysis should be considered. First, the heterogeneity of this meta-analysis was somewhat high. Second, we could not conduct some stratification analyses of other potential factors including smoking and drinking. Third, our results were based on unadjusted estimates for confounding factors, which might have affected the final results. Fourth, because of the lack of relevant data, potential gene–gene and gene–environment interactions were not performed. Fifth, the sample sizes of this meta-analysis were not large, which may lead to reduced statistical power. Last but not least, we did not explore the Asian populations due to lack of relevant studies.In conclusion, this meta-analysis indicates that Fas and FasL gene polymorphisms increase the risk of preeclampsia risk. Further studies with large sample sizes in other populations are urgently needed to confirm the findings of this meta-analysis.


Objectives: FAS plays a critical role in the extrinsic apoptosis pathway in autoimmune diseases. Previous studies investigating the association between FAS gene −670 A/G and −1377 G/A polymorphisms and the risk of autoimmune diseases reported controversial results. We performed the meta-analysis to evaluate the possible association. Methods: Relevant studies were identified by searching the PubMed, Embase, CNKI, and Wanfang databases up to December 2018. Odds ratios (ORs) and corresponding 95% confidence intervals (CIs) were calculated to determine the association. Results: A total of 43 articles including 67 studies (52 studies for FAS −670 A/G and 15 studies for −1377 G/A) were included in the meta-analysis. Our meta-analysis showed that the FAS −670 A/G polymorphism was associated with the risk of autoimmune diseases , especially in Caucasians , Asians , systemic lupus erythematosus (SLE) , multiple sclerosis (MS) , systemic sclerosis (SSc)  and Hashimoto’s thyroiditis (HT) ; the FAS −1377 G/A polymorphism was associated with the risk of autoimmune diseases , especially in Asians  and high quality studies . Conclusion: This meta-analysis demonstrated that the FAS –670A/G and –1377 G/A polymorphisms were associated with the risk of autoimmune diseases. Autoimmune diseases are chronic disorders characterized by the loss of immune tolerance to self-antigens, leading to immune-mediated tissue destruction. They affect 4–5% of adults, the majority of whom are women . Co-occuFAS gene is located on chromosome 10q24.1 in humans and is highly polymorphic [FAS promoter region [FAS −670 A/G and −1377 G/A polymorphisms may destroy signal transducer and activator of transcription protein 1 (STAT1) and stimulatory protein 1 (SP1) transcription factor binding sites, resulting in reduced promoter activity and FAS expression [FAS  is a cell surface receptor that belongs to the tumor necrosis factor (TNF) receptor superfamily . FAS is ymorphic . In somer region . The FASpression . Abnormapression .FAS −670 A/G rs1800682 and −1377 G/A rs2234767 polymorphisms and the risk of autoimmune diseases [FAS −670 A/G or −1377 G/A polymorphisms and some autoimmune diseases. However, these studies only analyzed SLE, RA, LN, SSc, pSS, JIA, SPA, and AIH and did not include all autoimmune diseases. Furthermore, previous meta-analyses [FAS −670 A/G or −1377 G/A polymorphisms is associated with autoimmune diseases risk by including 23 new articles, consisting of 33 studies [FAS polymorphisms with the risk of autoimmune diseases, including SLE, RA, MS, AIH, LN, SSc, AA, pSS, HT, GBS, PBC, vitiligo, GD, T1D, IAA, JIA, and SPA.Many studies have investigated the relationship between the diseases ,17,20–60diseases ,61–66 haanalyses ,65 incluanalyses ,30,31,40 studies ,59,60 onThis meta-analysis was conducted and reported according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) 2009 checklist .Literature published in English and Chinese was retrieved from the PubMed, Embase, CNKI, and Wanfang databases up to December 2018. The search strategy used the following medical subject heading (MeSH) terms combined with text words: ‘FAS or TNFRSF6 or CD95 or APO-1 or rs1800682 or rs2234767’, ‘polymorphism, genetic or polymorphisms or polymorphism or variant or mutation’ and ‘autoimmune diseases or autoimmune disease or autoimmunity’. A manual search of the reference lists was also performed to identify additional articles.FAS −670 A/G or −1377 G/A polymorphisms and autoimmune diseases risk; (2) available and sufficient genotype data to calculate the odds ratio (OR) with 95% confidence interval (CI); and (3) a case–control study design.Studies meeting all the following criteria were included in the analysis: (1) evaluation of the association between the Studies were excluded if they met the following criteria: (1) containing overlapping data; (2) not containing genotype data from the cases and controls; and (3) reviews, case reports, abstracts, letters, animal experiments and meta-analyses.P-value of test for Hardy–Weinberg equilibrium (HWE) in controls.Two investigators independently assessed and extracted data from all included studies. Discrepancies were resolved by discussion. The following data were collected from each study: disease type, first author, year of publication, country, ethnicity, genotyping method, sample sizes of cases and controls, genotype frequencies in cases and controls, and The methodological quality of the included studies was assessed independently by two investigators using the Newcastle–Ottawa scale (NOS) score . The NOSP<0.05 was considered to deviate from HWE. The ORs with their 95% CIs were used to assess the strength of associations between the FAS −670 A/G and −1377 G/A polymorphisms and autoimmune diseases. The statistical significance of the pooled ORs was determined by the Z test. The allelic , homozygous , heterozygous , dominant , and recessive  models were examined. The between-studies heterogeneity was assessed by Q test and quantified by I2 test [P≥0.1 or 2I < 50%, there was no heterogeneity, and pooled OR estimates were combined using the fixed-effects model (Mantel–Haenszel method); otherwise, the random-effects model (Mantel–Haenszel method) was used to combine summary data [P<0.05 were considered significant.The chi-square test was applied to examine whether the observed genotype frequencies in controls conformed to HWE, and  I2 test . When P≥ary data . If therary data  to evaluary data . All stahttp://www.ctu.dk/tsa/) in the allelic model with the overall included studies by setting an overall type I error of 5%, power of 80%, relative risk reduction (RRR) of 20%, and control event proportion [Traditional meta-analysis may yield type I errors due to dispersed data or repetitive significance testing when new studies are added to it ,75. Triaoportion . If the FAS −670 A/G polymorphism, and 15 studies examined the FAS −1377 G/A polymorphism. The characteristics of the articles included in the meta-analysis are summarized in A flowchart of the selection of eligible articles is presented in FAS −670 A/G and −1377 G/A polymorphisms and autoimmune diseases is shown in FAS −670 A/G polymorphism, a significant association between FAS −670 A/G and the risk of autoimmune diseases was observed under the heterozygous genetic model . In the FAS −1377 G/A polymorphism, our results indicated that FAS −1377 G/A polymorphism was associated with the risk of autoimmune diseases .A summary of the meta-analysis of the association between the Based on ethnicity, disease type, and quality score, we performed stratification analyses.FAS −670 A/G and −1377 G/A polymorphisms and autoimmune diseases risk stratified by ethnicity, disease type, and quality score are shown in The results of the meta-analysis of the association between the FAS −670 A/G polymorphism and the risk of autoimmune diseases in Caucasians  and Asians  but not in other ethnic groups. The association between FAS −1377 G/A polymorphism and the risk of autoimmune diseases was observed in Asians  but not in Caucasians.On the basis of ethnicity, the stratified meta-analysis showed an association between FAS −670 A/G polymorphism might be associated with the risk of SLE , MS , SSc , and HT . However, no association was observed between the FAS −670A/G polymorphism and the risk of RA, AIH, AA, pSS, GBS, PBC, or LN. For FAS −1377 G/A polymorphism, subgroup analysis was not performed owing to the limited study number.On the basis of disease type, the stratified meta-analysis suggested that the FAS −670 A/G polymorphism might not be associated with autoimmune diseases in high- or low-quality studies. However, the association between FAS −1377 G/A polymorphism and the risk of autoimmune diseases was observed in high-quality studies  but not in low-quality studies.On the basis of quality score, the stratified meta-analysis suggested that the FAS −670 A/G polymorphism and autoimmune diseases risk, and quality score may be the factor of heterogeneity across all studies of association between FAS −1377 G/A polymorphism and autoimmune diseases risk.Stratification analysis showed that ethnicity, disease type, and quality score might be the factors of heterogeneity across all studies of association between FAS −670 A/G polymorphism and SLE, RA, MS, AIH, LN, SSc, AA, and pSS are summarized in FAS −1377 G/A polymorphism, subgroup analysis was not performed owing to the limited study number). An association between the FAS −670 A/G polymorphism and the risk of autoimmune diseases was observed in Asian patients with SLE  or AIH  and in Caucasian patients with SLE , MS , or SSc . However, no significant risk was found in any specific ethnicity for RA, LN, AA, or pSS.The associations between the FAS −670 A/G polymorphism, the results from Egger’s tests indicated evidence for publication bias in the homozygous model for pSS, heterozygous models for Caucasians, AIH and high-quality studies, and recessive models for high-quality studies . After adjustment by the trim-and-fill method, the ORs corrected for publication bias were not qualitatively different for the five models . No publication bias was found among the studies regarding the association between Fas −1377 G/A polymorphism and autoimmune diseases risk . Therefore, the presence of publication bias did not influence the stability of the results. In addition, the results concerning association between FAS −670 A/G polymorphism and SLE, RA, MS, AIH, LN, SSc, AA, and pSS stratified by ethnicity did not show any evidence of publication bias  after excluding the HWE-deviating studies, but the association between the FAS −670 A/G polymorphism and autoimmune diseases risk was not qualitatively altered under the heterozygous model (P=0.038 vs. 0.006), illustrating that the meta-analysis results were stable. In the stratification analysis by ethnicity, the results in Caucasians and Asians did not change when the HWE-deviating studies were excluded. In the stratification analysis by disease type, the OR changed only under the recessive model (P=0.071 vs. 0.034) after excluding the HWE-deviating studies from the analysis of SLE, but the association between the FAS −670 A/G polymorphism and SLE risk was not qualitatively altered under the allelic model (P=0.001 vs. <0.001). However, the association between FAS −670 A/G and MS risk was materially altered under the dominant model (P=0.043 vs. 0.261) after excluding the HWE-deviating studies. Similarly, a change was observed in the analysis of Caucasian patients with MS under the dominant model (P=0.018 vs. 0.139). In the stratification analysis by quality score, the association between FAS −670 A/G and high-quality studies was materially altered under the heterozygous and dominant model  after excluding the HWE-deviating studies. Additionally, the results of the association between FAS −1377 G/A and autoimmune diseases risk did not change when the HWE-deviating studies were excluded in five models.The genotype frequencies in the controls of five articles ,51,52,55FAS −670 A/G: GG vs. GA; FAS −1377 G/A: A vs. G; FAS −1377 G/A: AA vs. GG; FAS −1377 G/A: AA+AG vs. GG) of the FAS −670 A/G and −1377 G/A polymorphisms were truly associated with an increased risk of autoimmune diseases . Furthermore, with regard to the FAS −670 A/G polymorphism, noteworthy results were found in Asians, Caucasians, SLE, HT, SSc, and MS. Regarding the FAS −1377 G/A polymorphism, a positive association was observed in Asians and high-quality studies.The FPRP values were calculated for the main significant associations and the results are shown in FAS −670 A/G polymorphism and autoimmune diseases risk, the cumulative Z-curve neither crossed conventional boundary nor trial sequential monitoring boundary, however, the sample size reached RIS (3365) in allelic model  and the cumulative Z-curve crossed the conventional boundary, although the cumulative Z-curve did not cross trial sequential monitoring boundary in allelic model  but was not associated with other autoimmune diseases. These findings may reflect differences in the risks of various autoimmune diseases due to differences in environmental and genetic backgrounds. The present results indicate that the FAS −670 G allele is associated with a decreased risk of SLE, MS, and AIH (in Asians), which conflicts with a previous finding that the FAS −670 G allele in the FAS promoter was associated with an increased risk of autoimmune diseases [FAS mRNA resulting in a variant that lacks exon 6, which encodes the transmembrane domain of FAS [FAS −670 G allele was associated with an increased risk of autoimmune diseases in Caucasians and with a decreased risk in Asians. For FAS −1377 G/A polymorphism, subgroup analysis was not performed owing to the limited study number. The FAS −1377 G/A polymorphism occurs at the consensus sequence of transcription factor SP1 binding site in the silencer region [FAS −1377 A allele may destroy SP1 transcription factor binding sites, resulting in reduced promoter activity and FAS expression [In the disease-type stratification analysis, the diseases ,78. One diseases ,80. Trandiseases , with aln of FAS . Plasma n of FAS ,82–85. Sr region . The FASpression . Abnormapression .FAS −670 A/G polymorphism might be associated with the risk of rheumatic disease, especially in Asians, SLE and RA, and the FAS −1377 G/A polymorphism was associated with SLE risk. Compared with this meta-analysis, our meta-analysis focused on overall autoimmune diseases risk and showed that FAS −670 A/G polymorphism was associated with autoimmune diseases risk in Caucasians, MS, SSc and HT; and the FAS −1377 G/A polymorphism was associated with autoimmune diseases risk in Asians and high quality studies, which were different from the previous meta-analyses. One meta-analysis [FAS −670 A/G polymorphism may be associated with SLE risk in the Chinese population. Two meta-analyses [FAS −670 A/G and −1377 G/A polymorphisms was associated with the risk of SLE, stratification by ethnicity indicated an association between the FAS −670 A/G and SLE in Asian populations. Two meta-analyses [FAS −670 A/G polymorphism was not associated with the risk of RA. One meta-analysis [FAS −670 A/G polymorphism was not associated with the risk of AIH. These six meta-analyses focused on the association between FAS polymorphism and a single disease . Compared with these meta-analyses, our meta-analysis covered overall autoimmune diseases, and subgroup analyses were performed by ethnicity, disease type, and quality score, thereby yielding several novel and distinct findings. Furthermore, some previous meta-analyses [FAS polymorphism and autoimmune diseases.We performed a meta-analysis of data from patients diagnosed with autoimmune diseases  and healthy controls. This meta-analysis differs from the seven previous meta-analyses ,61–66 beanalysis  showed tanalyses ,66 suggeanalyses  showed tanalysis  suggesteanalyses ,65 incluanalyses ,30,31,40analyses ,59,60 onSensitivity analysis are generally performed to assess the robustness of meta-analyses by excluding and including HWE-deviating studies from genetic association studies, which is a recommended approach . ProbablFAS gene and did not investigate associations between genetic haplotypes containing the FAS −670 A/G and −1377 G/A polymorphisms and the risk of autoimmune diseases because of inadequate haplotype data. It is unknown whether other genetic mutations contribute to changes in the expression or function of the FAS gene. For uncovering the genetic causes of disease, haplotypes provide more information and have a greater influence than genotypes and single SNPs. Fourth, most studies included in our analysis were performed in the Caucasian and Asian populations; therefore, our results may apply only to these ethnic groups. Additional studies of other ethnicities are needed. Fifth, autoimmune diseases are multifactorial diseases caused by interactions between genetic and environmental factors, meaning that the FAS −670 A/G and −1377 G/A polymorphisms may only partially influence the pathogenesis of autoimmune diseases; this may lead to bias in the present results. Finally, the findings of our meta-analysis should be interpreted with caution in the case of heterogeneity observed under some genetic models.The present study has several limitations that should be considered when interpreting the conclusions. First, only case–control studies were considered for inclusion. Selection bias and unmeasured confounding can occur at both the design and analysis stages of observational studies. Second, this analysis only included articles published in English and Chinese; this may reduce the credibility of the results because of language bias . Third, Translating information of genetic associations into clinical diagnostics would help with improved understanding of the autoimmune diseases’ etiology. Establishing evidence-based medical evidence of genetic susceptibility to autoimmune diseases risk might facilitate the preventive and therapeutic strategies, which has a beneficial clinical utility for not only clinicians and researchers but also patients.FAS −670 A/G polymorphism might be associated with the risk of autoimmune diseases, especially in Caucasians and Asians, SLE, MS, SSc, and HT. Moreover, the FAS −670 A/G polymorphism might be associated with the risk of autoimmune diseases in Asian patients with SLE or AIH and Caucasian patients with SLE, MS, or SSc. The FAS −1377 G/A/ polymorphism might be associated with the risk of autoimmune diseases, specifically for Asians and high quality studies. Stratification analysis showed that ethnicity, disease type and quality score might be the factors of heterogeneity across all studies of association between FAS −670 A/G polymorphism and autoimmune diseases risk, and quality score might be the factor of heterogeneity across all studies of association between FAS −1377 G/A polymorphism and autoimmune diseases risk.In summary, our meta-analysis suggested that the


I2 = 0%, P = 0.44), 0.96 (0.82–1.12) for fruits without low heterogeneity  and 0.94 (0.84–1.06) for vegetables with low heterogeneity . Dose-response analyses also showed no significantly inverse associations for each 100 g/day increase; the summary relative risks were 1.00 (0.98–1.02) for total fruits and vegetables, 1.01 (0.97–1.05) for fruits and 1.00 (0.97–1.03) for vegetables. The results of subtype analyses were consistent with the fruit and vegetable analyses; the relative risks were 0.97 (0.80–1.17) for citrus fruit without low heterogeneity  and 0.89 (0.76–1.05) for cruciferous vegetables without low heterogeneity . In conclusion, this meta-analysis does not support significant associations between fruit and vegetable consumption and pancreatic cancer risk.The associations between fruit and vegetable consumption and pancreatic cancer risk are inconclusive. We conducted a meta-analysis of prospective studies to investigate the associations. The search was conducted systemically using the PubMed and EMBASE databases up to March 2017. Relative risks and 95% confidence intervals for the highest versus lowest consumption and dose-response analyses were assessed. Subtype and subgroup analyses were performed. Twelve studies were eligible. The summary relative risks of the highest versus lowest consumption were 0.95 (0.80–1.12) for total fruits and vegetables without heterogeneity ( Pancreatic cancer (PC) is a highly sinister disease with an extremely poor prognosis and a five-year survival rate of less than 5% . PC is tTherefore, in consideration of the large burden of PC worldwide and the controversial evidence, we conducted an updated meta-analysis of prospective studies based on a quantitative amalgamation of the eligible data with the following objectives: (1) to gain a better understanding of the associations of fruit and vegetable consumption with PC risk; (2) to further examine the associations according to subtype analyses for citrus fruit and cruciferous vegetables and subgroup analyses for gender and other factors, including geographic area, sample size, publication year, periods of follow-up and main adjustments; and (3) to further evaluate the dose-response associations between fruit and vegetable consumption and PC risk.Figure P = 0.31) without heterogeneity  . We further checked for nonlinearity of the dose-response relationship between fruit and vegetable consumption and PC risk, and there was no evidence of a potential nonlinear relationship (Pnonlinearity = 0.56).Five studies were included, and the RR per 100 g/d increase in fruit and vegetable consumption was 1.00 (0.98–1.02) without heterogeneity  of the included studies. Subgroup analyses showed that the differences in the RRs were not significant for the geographic area, sample size, publication year, periods of follow-up and all of the adjustable variables (Table There was no significant heterogeneity (es Table .P = 0.511) and Begg’s test (P = 0.452) did not suggest significant evidence of publication bias. The sensitivity analysis suggested that the change in recalculated RRs was not significant, with a range from 0.87 (0.73–1.05) when excluding Inoue-Choi 2011 to 0.95 (0.80–1.12) when excluding Shigihara 2014.The funnel plot , Egger’sP = 0.29, I2 = 12%) and in women  with heterogeneity .Two studies with men and 2 studies with women were included. The results Table  indicateP = 0.57) without heterogeneity  . We further checked for nonlinearity of the dose-response relationship, and there was no evidence of a potential nonlinear relationship between fruit consumption and PC risk (Pnonlinearity = 0.15).Eight studies were eligible, and the RR per 100 g/d increase in fruit consumption was 1.01 (0.97–1.05) without heterogeneity  of the included studies. Subgroup analyses showed that the differences in the RRs were not significant for the geographic area, sample size, publication year, periods of follow-up and all of the adjustable variables except for alcohol  (Table There was no significant heterogeneity (P = 0.911) and Begg’s test (P = 0.917) did not suggest significant evidence of publication bias. The sensitivity analysis suggested that the change in recalculated RRs was not significant, with a range from 0.89 (0.78–1.01) when excluding Nothling 2007 to 1.02 (0.89–1.08) when excluding George 2009.The funnel plot , Egger’sP = 0.21, I2 = 35%). An RR of 0.80 (0.66–0.98) suggested a significant association between fruit consumption and PC risk in men without heterogeneity .Three studies for men and 3 studies for women were included. The results Table  indicateP = 0.15, I2 = 39%)  and Begg’s test (P = 0.133) suggested no significant evidence of publication bias.Six studies were eligible and, the RR was 0.97 (0.80–1.17) without significant heterogeneity  Figure . The sen) Figure , Egger’sP = 0.32) without heterogeneity   Figure , Table 3P = 0.40, I2 = 4%) .Eight studies were eligible, and the RR per 100 g/d increase in vegetable consumption was 1.00 (0.97–1.03) without heterogeneity  Figure . We furtP = 0.36, I2 = 9%) of the included studies. Subgroup analyses showed that the differences in the RRs were not significant for geographic area, sample size, publication year, periods of follow-up and all of the adjustable variables (Table There was no heterogeneity (P = 0.649) and Begg’s test (P = 0.721) did not suggest significant evidence of publication bias. The sensitivity analysis suggested that the change in the recalculated RRs was not significant, with a range from 0.91 (0.81–1.03) when excluding Heinen 2011 to 0.98 (0.87–1.10) when excluding Inoue 2003.The funnel plot , Egger’sP = 0.28, I2 = 22%) and in women  without heterogeneity .Four studies for men and 4 studies for women were included. The results Table  indicateP = 0.32, I2 = 14%)  and Begg’s test (P = 0.452) suggested no significant evidence of publication bias.Six studies were eligible, and the RR was 0.89 (0.76–1.05) without heterogeneity  Figure . The sen) Figure , Egger’sIn general, fruit and vegetable consumption has been reported to be protective against diabetes , cardiovOur findings provide more detailed evidence that high total consumption of fruits and vegetables, fruits, vegetables, citrus fruit or cruciferous vegetables is not associated with a decreased risk of PC overall; although increased consumption of fruits but not vegetables is associated with a lower PC risk in men but not in women, the included studies are limited. Furthermore, dose-response analyses suggested that there are no significant dose-response relationships between a 100 g/d increment in fruit and vegetable consumption and PC risk. Additionally, the results of subgroup analyses and subtype analyses were consistent with the original analyses. Overall, our analyses based on prospective studies showed that there was no evidence of associations between fruit and vegetable consumption and pancreatic cancer risk, and the detailed findings including subgroup analyses and subtype analyses further clarify the associations between fruit and vegetable consumption and PC risk and can be used as a reference for the update of dietary guidelines.Our study had several strengths. The first strength is that a long duration of the follow-up and the large sample size of the included studies provided robust evidence to date and increased the statistical power. Second, included studies were identified from 13 countries in Europe, North America and Asia, which increased the statistical generalizability. Third, detailed subgroup analyses were conducted according to the main potential confounders of the studies and main adjustable variables of PC, including gender. These independent results increased the significant power and provided more detailed data of reference significance for dietary guidelines concerning PC worldwide. Fourth, we performed dose-response analyses in addition to simply performing comparisons of the lowest versus highest categories, which further verified our results. Fifth, we conducted further subtype analyses for the two main types of fruits and vegetables reported: citrus fruit and cruciferous vegetables; these analyses strengthened our findings in the details. Finally, the heterogeneity between the studies was not statistically significant in all of the analyses except the analysis of fruits and vegetables in women. Additionally, publication bias was negligible based on the results of the funnel plots, Egger’s tests, Begg’s tests and sensitivity analyses. These statistical results increased reliability of our data.Several limitations of this study must be taken into consideration. First, the included studies were observational studies, which were liable to residual confounding and other unmeasured factors. Nevertheless, the main confounders were adjusted in most included studies, and we further conducted subgroup analyses to assess the effects of these confounders, including sex, smoking, alcohol use, body mass index, family history of PC and history of diabetes mellitus. In general, these findings were similar to the summary estimates and were consistent for each of the subgroup analyses except for alcohol and fruit consumption. Nevertheless, storage conditions, production methods, cooking methods and nutrient content might be different among the included studies, and measurement errors to assess dietary consumption can lead to bias; and we cannot thoroughly exclude potential residual confounding.Second, the exposure ranges from the lowest to highest categories were different among the included studies, which contributed to incomparable results and heterogeneity to some extent. However, we adopted and pooled the RRs for the comparison of the highest versus lowest category, and the dose-response analyses verified the results. Additionally, there was no statistically significant heterogeneity in all of the analyses except the analysis of fruits and vegetables in women.Third, although we further explored the associations between the consumption of citrus fruit or cruciferous vegetables and PC risk, there were no subtype analyses for other types of fruits and vegetables. Therefore, our findings should not be used to analyze other specified fruits or vegetables. Additionally, because we only focused on PC, the available results should not be used to determine associations with pancreatic intraepithelial neoplasia and pancreatic benign tumors.Last but not least, the quality of two included studies was not high , 16 and Last but not least, the dietary information of all of the included studies was limited to middle-aged and older persons. The average ages were more than 50 years old in all of the studies, and most of the subjects were older than 60 years of age. Our findings cannot capture the pertinent exposure period of whether fruit and vegetable consumption during childhood, adolescence or early adulthood may be protective against PC.Prospective cohort studies and nested case-control studies were included.Histological features that were not consistent with the diagnostic gold standard of PC such as pancreatic benign tumors and pancreatitis were excluded.Systematic reviews, meta-analyses, narrative reviews, pooled analyses were excluded.Comments, editorials, letters, case reports were excluded.Studies in which only the abstract could be obtained were excluded.Data that could not be combined or that were incomplete were excluded.Studies were limited to those involving humans; and the publication language of the included studies was limited to English.The highest quality studies, the largest sample sizes and the most recent studies when identifying studies with the same patient cohort were selected.“fruit”, “fruits”, “vegetable”, “vegetables”, “diet”, “dietary”, “food” and “foods” in combination with “gastrointestinal/digestive/alimentary/pancreatic/pancreas”. We identified additional literature by manually searching the reference lists of the reviews and the extracted studies. The two sets of keywords were combined individually. The eligibility criteria were judged independently by two authors .We comprehensively identified eligible studies through searching the PubMed and EMBASE databases up to March 2017 using the following search terms: Two researchers  independently assessed the study quality using the Newcastle-Ottawa Scale (NOS) . The NOSTwo researchers  independently extracted primary relevant data from the studies. Disagreement was resolved by reaching a consensus. The first author, country, year of publication, study period, study population, method of dietary assessment, dietary exposure categories, type of dietary exposure measured, adjusted RR (95% CI) (highest to lowest), adjustments and NOS score of each study were summarized and shown in a data extraction sheet Table . For eacPnonlinearity < 0.05 indicates a nonlinear model.Random-effects models were used to quantify the associations between fruit and vegetable consumption and PC risk. The method described by Greenland and Longnecker  was usedQ and I2 statistics according to the updated version 5.1.0 of Cochrane Handbook for Systematic Reviews of Interventions . Subgroup analyses were conducted to further explore the sources of heterogeneity by publication year, geographic area, sample size, period of follow-up and adjustments .Heterogeneity among the studies was assessed using Cochran P < 0.1 was considered to indicate significant publication bias) [We used funnel plots, Begg’s test and Egger’s test to assess publication bias (on bias) . We condData were collected and extracted using SPSS 17.0 . STATA version 12.1  and RevMan5.3  were used for synthesis and analysis, respectively.This meta-analysis does not support significant associations between fruit and vegetable consumption and pancreatic cancer risk. The observed decrease in PC risk with fruit consumption in men needs further investigation because of limited included studies.


A recent meta-analysis compared the relative risks of colorectal cancer between the highest and lowest levels of coffee consumption in the Japanese population. However, this analysis did not define the risks with respect to specific exposure values when considering levels of coffee consumption per day in the study population.We conducted a two-stage random-effects dose-response meta-analysis of the association between coffee consumption and colorectal cancer among the Japanese. This was performed by modeling coffee consumption using restricted cubic splines to be able to examine a potential nonlinear relation.We identified a total of 26 studies from seven articles, which were distributed separately according to sex and colon/rectum cancers. Data from 14 cohort studies showed that the pooled relative risks for colorectal cancers were less than 1.0 in cases with coffee consumption of 1–3 cups/day and 1.0 in cases with consumption of 4 cups/day or more, although these results were not statistically significant. Data from 12 case-control studies showed that the pooled odds ratios for cancer risk were significantly less than 1.0 in cases with coffee consumption of 1–6 cups/day.Findings from this meta-analysis indicate that moderate coffee consumption may not be associated or may be weakly inversely associated with the risk of colorectal cancer in the Japanese population. One of the reasons for this might be due to the differences in the areas/countries studied.These conclusions are supported by a number of studies, including prospective cohort and population-based case-control studies, as well as meta-analyses. Although several meta-analyses16 This study concluded that the findings of cohort studies were inconsistent, as they reported a relative risk (RR) of 0.95 with 95% confidence interval (CI) of 0.77–1.17, whereas findings of case-control studies suggested an inverse association, reporting an odds ratio (OR) of 0.78 with 95% CI of 0.65–0.95. Akter et al thus concluded that the evidence was insufficient to support the fact that coffee drinking either increases or decreases colorectal cancer risk among the Japanese population.Recently, a meta-analysis of epidemiological studies on coffee consumption and CRC risk conducted only for the Japanese population was reported by Akter et al.17 Moreover, it would be of great interest to know the risks with respect to specific exposure values when considering coffee consumption per day.Most of the previous meta-analyses were performed using the pooled analysis of RRs or ORs as reported in each cohort and case-control study. The aforementioned meta-analysis among the Japanese population evaluated CRC risks for the highest coffee-consumption group as compared with those of the lowest. However, categories of coffee consumption per day differed between studies, and study-specific risks of the highest consumption categories relative to the lowest were adopted in the analyses. Thus, interpretation of the pooled results across study populations might have been complicated when different categories were used.17,18 have presented dose-response meta-analyses of worldwide populations by employing a cubic spline model to examine a potential nonlinear (including a simple linear) association between coffee consumption and CRC. However, this type of analysis had not been performed among the Japanese population. Therefore, we conducted such a dose-response meta-analysis of the association between coffee consumption and CRC in this population alone, based on the studies pooled in the recent meta-analysis by Akter et al.16A dose-response meta-analysis could provide a solution to this problem. Some recent studies16—five cohort studies and nine case-control studies—along with the most recently published19 case-control study. Only studies based on Japanese residents were included. We employed articles reporting RRs or ORs along with details, such as the number of people, cases, standard errors, or CIs, in more than three categories.We evaluated articles of relevant epidemiological studies identified by Akter et al17,18 of the association between coffee consumption and CRC (Appendix). We assessed dose-response associations using a summarized group of data that was based on the assignment of levels to grouped exposure intervals by applying the likelihood-based assignment procedure.20,21 We employed the restricted cubic spline model for potential nonlinear dose-response curves for selecting locations of the three knots appropriately in order to maximize the likelihood.21,22 This procedure can estimate the log-RR with respect to the exposure level of zero for each study, even when the level of reference category for reported RRs or ORs is not zero. Statistical heterogeneity among studies was assessed using the multivariate Cochran’s Q-test. Additionally, goodness of fit assessments were conducted by indicating as deviance.23 These analyses were performed using the package “mvmeta”24 in software R, version 3.3.1 .25We conducted a two-stage random-effects dose-response meta-analysisWe performed dose-response meta-analyses for cohort studies and case-control studies separately. For sub-group analyses, we performed combined analyses among studies of colon and rectum cancers, respectively. In addition, we performed separate analyses for each gender in the cohort studies. However, we were unable to perform these analyses for case-control studies, because only one study reported separate results by gender.19,26–31 separately. Among the cohort studies, three articles27–29 separated results of studies for cancers of the colon and rectum, and one article26 reported only on colon cancer. All of these articles reported results of studies for men and women separately. A total of 14 cohort studies were included in this analysis  from four cohort articles and three case-control articles,1, β2) =  and three knots  =  , than the simple linear model . The pooled RRs of total CRC were less than 1.0 for coffee drinkers of 1–3 cups/day and >1.0 for 4 cups/day or more consumed, as compared to subjects with no coffee consumption; however, these results were not significant, with α = 0.05 for all levels of coffee consumption.For the overall association between coffee consumption and total CRC , we were able to estimate a pooled dose-response curve from the 14 cohort studies using a cubic spline regression method with coefficients  =  with  =   for the cubic spline model showed a better fit than the simple linear model . The pooled ORs of total CRC were significantly less than 1.0 for coffee drinkers of 1–6 cups/day as compared to subjects with no coffee consumption.On the other hand, a pooled curve from the 12 case-control studies had coefficients (βQP = 0.351 (colon) and 0.843 (rectum) in cohort studies, while these values in case-control studies were 0.155 (colon) and 0.117 (rectum).Figure Table 16 which only compared RR or OR between the highest consumption group and the lowest. Our dose-response meta-analysis indicated the CRC risk by specific levels of consumption, and results suggested that mild coffee consumption may not be associated or may be weakly inversely associated with risk of CRC.In this study, a meta-analysis of cohort studies indicated that associations between coffee consumption and CRC were not significant, whereas meta-analysis of case-control studies suggested an inverse association in the Japanese population. These findings coincide with those of the report by Akter et al,16 this finding is generally consistent with previous meta-analyses of case-control studies,9,14,15 including Japanese studies. In contrast to Akter’s analysis, we employed studies from three articles, including a new report, based on which we excluded five articles. Although this is a limitation of our analysis, it should be noted that all articles excluded were older reports, published from 1984 through 1997, and included articles that were published after 1998. This suggests that the same risk reduction might also be found in a recent study of the Japanese population. Therefore, due to limitations inherent in case-control studies, findings from these studies should be interpreted with caution. Moreover, differences in preferences of coffee consumption between populations where cases and controls were selected could raise the possibility of selection bias.Case-control studies, in particular those employing summarized ORs, have reported a statistically significant decrease in risk of CRC with levels of coffee consumption of 1–6 cups/day. Moreover, they showed an inverse association with both colon and rectum cancers. As pointed out by Akter et al,In cohort studies, on the other hand, we should be cautious because we cannot be sure that coffee-drinking habits did not change after baseline. Moreover, we should be deeply concerned about whether it was common to drink coffee in the early 1990s in Japan, as well as what the characteristics of the population with coffee habits were at that time. The cohort-study populations were generally older in age than those of the case-control studies, even if, for example, both studies enrolled 40-year-olds, because the age of the entry was recorded in the cohort study. Considering this older population, particularly in men, prevalence of tobacco consumption and/or exposure might be high, thus presenting a confounding effect that could not be sufficiently adjusted for. Moreover, as compared to the colon, the rectum is the most affected by smoking, and, in the cohort studies, only cancer of the rectum showed an increased association between coffee consumption and CRC.22,32,33 Therefore, further discussion of these models, including evaluation of different methods, is warranted.As a strength of this study, our dose-response meta-analysis reveals the CRC risk in specific levels of coffee consumption, rather than the relative risks between the highest and lowest levels of consumption provided in traditional meta-analysis. Our meta-analysis assessed potential nonlinear associations between coffee consumption and the risk of CRC, employing the restricted cubic spline model. On the other hand, as a limitation of this study, the results of a meta-analysis must be sensitive to the model, including the number of knots, although some studies have applied similar models to evaluate the dose-response of coffee consumption and disease.From the results of the present dose-response meta-analysis, we conclude that there is insufficient evidence to support the hypothesis that coffee drinking either increases or decreases CRC risk among the Japanese population. Moreover, our results suggest that mild coffee consumption may not be associated with or only weakly inversely associated with the risk of CRC, although its influence is not statistically significant.


Alteration in serum expression of Transforming Growth Factor-beta (TGF-β) and IL-10 have been suggested to play a role in the pathogenesis of Kawasaki Disease (KD). Inconsistent reports exist on the association of IL-10 polymorphisms with KD susceptibility and Coronary Artery Aneurysms (CAA).A number of 110 paediatric patients with KD and 140 healthy individuals were recruited to investigate the frequency of Single Nucleotide Polymorphisms (SNPs) of TGF-β C/T at codon 10 (rs1982073), C/G at codon 25 (rs1800471) and IL-10 A/G at −1082 (rs1800896), C/T at −819 (rs1800871) and A/C at −592 (rs1800872) and their respective genotype and haplotypes. A comprehensive search was performed in MEDLINE and SCOPUS using the keywords of interleukin 10, transforming growth factor beta, and Kawasaki disease. Moreover, previous studies investigating the TGF-β and IL-10 polymorphisms in KD were evaluated. Review Manager Version 5.1 Software was used to perform meta-analysis.vs. 1471 healthy controls showed no difference in the frequency of any of the IL-10 genetic variants in KD patients, regardless of the presence of CAA.There was no significant association between allelic or genotypic variants in the mentioned polymorphisms in TGF-β or IL-10 with KD or CAA. The only significant haplotypic variant was TC variant at codon 10, and 25 of TGF-β polymorphisms were associated with higher risk of KD. Meta-analysis of a total number of 770 patients Polymorphisms of TGF-β or IL-10 are not associated with additional risk for KD in Iranian population. IL-10 polymorphisms at −1082, −819 and −592 positions are not associated with KD, nor do they predict coronary artery aneurysm formation. Kawasaki Disease (KD) also known as mucocutaneous lymph node syndrome is an acute systemic vasculitis with strong predilection for coronary arteries During the initial acute stage of KD, markers of inflammation were abundantly found in plasma, helper CD4+ T-cells which dominate the peripheral blood Expression levels of FOXP3, which is a crucial transcription factor of Treg and Th17 cells is suppressed in acute febrile phase of KD, and surges to a peak three weeks after IVIG therapy TGF-β gene polymorphism itself. Single nucleotide variants of proinflammatory cytokines TGF-β (codon 10 and codon 25) and IL-10 genes were addressed. Considering the discrepancy in results of various studies on IL-10 polymorphisms including A/G at −1082, C/T at −819, and A/C at −592, meta-analytic evidence on association of IL-10 allelic and genotypic variants with KD susceptibility was provided.A large body of literature support association of polymorphisms of TGF-β signalling pathway with predilection to KD and CAA β and Zα are standard normal variates for power and level of significance, respectively. Adopting a statistical power of 80% (Zβ=0.84) and significance threshold of 95% (Zα/2=1.96), SD and mean differences reported by Hsieh et alet al’s statistics In this study, 110 Iranian patients  were recruited with acute KD, who were randomly selected from referrals to the Children’s Medical Center, the Pediatrics Center of Excellence in Tehran, Iran, and a group of 140 age-sex matched healthy individuals as the control group was selected. It was a single center prospective observational case-control study. Sample size was calculated based on the following formula The diagnosis of KD was primarily made according to American Heart Association Criteria This study was approved by local ethics committee and institutional review board of Tehran University of Medical Sciences in April 2017. Signed informed consent forms were obtained from parents of all enrolled cases to perform sampling and echocardiography when necessary, and for publication of the results as papers.ml of whole blood was taken from each participant and kept with EDTA (Ethylene-diamine-tetraacetic acid) until investigation . Genomic DNA was extracted from whole blood, using Phenol-Chloroform method Five The analyses were all performed using the Epi Info statistical software . Allele, genotype and haplotype frequencies were estimated by direct gene counting. All of the allele frequencies were in line with the Hardy-Weinberg equilibrium. Frequencies were analyzed using chi-square test or Fisher’s exact test and odds-ratios of 95% Confidence Interval (95%CI). All tests were two-sided and the probability of less than 0.05 was considered statistically significant.IL-10 genes and TGF-β gene polymorphisms in KD patients were included in English language, with no limitation on publication date or status. A comprehensive literature search through PubMed and Scopus was done by using the key words of interleukin 10, transforming growth factor beta, IL-10 OR TGF-β, KD, and mucocutaneous lymph node syndrome. The search was repeated and updated accordingly to the date of submission (August 2018). To identify additional studies, reference lists and contacted corresponding authors of all the publications included in the present systematic review and meta-analysis were scanned.For the meta-analysis, original articles investigating frequency of Two investigators independently reviewed results of 91 studies yielded through literature search according to preliminary inclusion criteria listed above, based on title and abstract. Any disagreement between reviewers was resolved by discussion or referral to a third investigator. Additional inclusion criteria were: 1) diagnosis of KD established by a pediatric rheumatologic disease and/or according to international criteria, 2) availability of either number or odds ratio or relative risk of the alleles, genotypes or haplotypes, 3) primary outcome of the study being KD, CAA or both. Studies on atypical KD, other complications of KD such as heart block or myocardial infarction, and those in which the control group was selected from related or non-related individuals with other cardiac complications, were excluded. Reference stratification and systematic reviews were performed using the SM Software. The following data were extracted from included publications. The data comprised the name of author, date of publications, mean age of case and control group, the study type of comparison/contrast, and allelic and genotype frequencies. The flow diagram of literature review and study selection is shown in As previous investigations on SNPs of TGF-β at codon 10 and 25 were scant, further analyses on these studies  were not done.Quality assessment was performed using the Newcastle-Ottawa Scale (NOS), for observational case-control studies 2I index. According to the Cochrane guidelines, the 2I less than 40% means that the inconsistency across studies is not important. In this case, fixed effects model was used. If the 2I estimates fluctuated more than 40%, the random effects procedure was used as the analysis model. A p-value less than 0.05 was considered statistically significant.Meta-analysis was performed using Review Manager , on the above selected five studies In the first part of the study, a matched-case-control research on Iranian patients with KD  and a group of age-sex matched healthy controls  was done. Allele frequencies of SNPs of TGF-β, IL-10 in the two groups are presented in vs. 1331 Healthy Controls (HC), added to 110 patients and 140 controls from our study, yielding a total number of 770 KD patients and 1471 HC for the meta-analytic population. Mean age of patients ranged from 1.25 to 3.29±2.36 years and Taiwan studies were divided into those comparing frequency of polymorphisms in KD patients vs. healthy controls, including results from current study, and those comparing polymorphisms in KD patients with CAA and those without CAA. One study had only addressed polymorphic variants at −592 (C/T) (33). Also, Weng et al reported results of their study in Thai population in two separate publications vs. HC and KD with CAA vs. KD without CAA). Total number of subjects in pooled analysis for KD vs. HC comparison was 572 patients and 776 HC, and 272 vs. 726 for KD+ CAA vs. KD-CAA contrast.Review of literature yielded five studies, investigating association of IL-10 polymorphisms in KD patient. Total number of patients included from previous studies were 660 patients vs. healthy individuals. The same result was attributed to subgroup meta-analysis of KD+CAA vs. KD without CAA , Wang et al reported higher frequency of C allele and CC gene in KD patients with coronary artery lesions in acute but not chronic stage KD et alWhile Hsueh TGF-β is able to induce vascular wall myofibroblast generation Although literature on TGF-β polymorphisms in KD patients is scant, polymorphisms in components of the TGF-β/SMAD signalling pathway greatly influence susceptibility to KD Common SNPs of TGF-β and IL-10 in patients with KD were investigated in our sample of Iranian population, which revealed no significant results in any of the allelic, genotypic or haplotypic components. Previous studies had reported conflicting results on association of IL-10 polymorphisms at −1082, −819, or −592 and TGF-β polymorphisms at codon 10 and codon 25, with KD and/or coronary artery aneurysm as its most lethal complication. Through an updated meta-analysis, no association was found between the mentioned SNPs and risk of KD or CAA. Although previous reports have suggested a dysregulation in anti-inflammatory cytokines as a mechanism for CAA formation, but these results are not supported by our meta-analysis results on common polymorphisms of TGF-β and IL-10. One major limitation of the study to be addressed is that studies that met our inclusion criteria had all taken place in Taiwan and this might limit extrapolation of the results.


O-6-methylguanine-DNA methyltransferase (MGMT) gene in the occurrence of cancer. Based on the available data from the online database, we performed an updated meta-analysis. We retrieved 537 articles from our database research and finally selected a total of 54 case–control studies (21010 cases and 34018 controls) for a series of pooling analyses. We observed an enhanced risk in cancer cases compared with controls, using the genetic models T/T compared with C/C  = 1.29) and T/T compared with C/C+C/T . We detected similar positive results in the subgroups ‘Caucasian’, and ‘glioma’ . However, we detected negative results in our analyses of most of the other subgroups (P>0.05). Begg’s and Egger’s tests indicated that the results were free of potential publication bias, and sensitivity analysis suggested the stability of the pooling results. In summary, the T/T genotype of MGMT rs12917 is likely to be linked to an enhanced susceptibility to cancer overall, especially glioma, in the Caucasian population.In the present study, we aimed at determining the potential role of rs12917 polymorphism of the  O-6-methylguanine-DNA methyltransferase (MGMT) protein, encoded by the MGMT gene located on chromosome 10 (10q26) . . O-6-met2917 Leu8Phe and re143Val) . In addiMGMT rs12917 polymorphism on its susceptibility to cancer, but the results were inconclusive. Before 2013, only three relative meta-analyses investigated the potential role of this polymorphism in the overall risk for cancer ) OR Methylated-DNA-Protein-Cysteine S-Methyltransferase) OR Methylated DNA Protein Cysteine S Methyltransferase) OR S-Methyltransferase, Methylated-DNA-Protein- Cysteine) OR O(6)-Methylguanine Methyltransferase) OR O(6)-Alkylguanine-DNA Alkyltransferase) OR O(6)-MeG-DNA Methyltransferase) OR O(6)-Methylguanine DNA Transmethylase) OR Guanine-O(6)-Alkyltransferase) OR O(6)-AGT) OR DNA Repair Methyltransferase II) OR DNA Repair Methyltransferase I) OR MGMT)) AND  OR Polymorphisms, Genetic) OR Genetic Polymorphisms) OR Genetic Polymorphism) OR Polymorphism (Genetics)) OR Polymorphisms (Genetics)) OR Polymorphism) OR Polymorphisms)) AND ((((((((((((((((((Neoplasms [MeSH Terms]) OR Neoplasia) OR Neoplasias) OR Neoplasm) OR Tumors) OR Tumor) OR Cancer) OR Cancers) OR Malignant Neoplasms) OR Malignant Neoplasm) OR Neoplasm, Malignant) OR Neoplasms, Malignant) OR Malignancy) OR Malignancies) OR Benign Neoplasms) OR Neoplasms, Benign) OR Benign Neoplasm) OR Neoplasm, Benign).P-value of Hardy–Weinberg equilibrium (HWE) <0.05; (O), lack of full genotype frequency data; (S), review, meta, poster, or meeting abstract. Eligible articles had to be designed as case–control studies, targetting the genetic relationship between MGMT rs12917 and cancer risk and containing the full genotype  frequencies in both cancer cases and negative controls.We designed our inclusion and exclusion criteria according to Patient, Intervention, Comparison and Outcome and Study design (PICOS) principles. We ruled out duplicates and screened improper articles. Exclusion criteria were as follows: (P), non-cancer patients; (I), other variants, gene expression or methylation; (C), lack of study controls or After extracting usable data, we listed the basic information in tables. We assessed methodological quality via the Newcastle–Ottawa Scale (NOS) . High-qu2 test. A P-value of Cochran’s Q statistic < 0.1 or 2I value > 50% was considered to show a high level of heterogeneity. We thus used the DerSimonian–Laird association test with a random-effects model. Otherwise, we used the Mantel–Haenszel association test with a fixed-effects model. The P-value of association test, summary odds ratio (OR), along with the corresponding 95% confidence interval (CI) could be obtained for the allele (T compared with C), homozygous (T/T compared with C/C), recessive (T/T compared with C/C+C/T), heterozygous (C/T compared with C/C), dominant (C/T+T/T compared with C/C), and carrier (T compared with C) models.We used STATA software version 12.0-SE  to perform our analyses. We first assessed the inter-study heterogeneity using Cochran’s Q statistic and the IWe performed subgroup analyses by race, cancer type, and control source. Additionally, we assessed possible publication bias by means of Begg’s and Egger’s tests and evaluated the robustness of the results through sensitivity analysis.MGMT rs12917 polymorphism and cancer risk via an overall meta-analysis. As shown in P-values of heterogeneity > 0.1; 2I < 50%). For other models , we performed a DerSimonian–Laird association test with a random-effects model. Pooling data  and T/T compared with C/C+C/T  genetic models. Nevertheless, we failed to detect any statistical difference between cancer cases and negative controls under other genetic models (P>0.05). Forest plot data are shown in MGMT rs12917 polymorphism was likely to be associated with an increased susceptibility to cancer.First, we studied the association between the ing data  indicateP<0.001; OR = 1.35), ‘glioma’ , ‘population-based control (PB)’  and ‘hospital-based control (HB)’ . Next, we carried out four subgroup analyses by race, cancer type, and control source. For the T/T compared with C/C model , the assP<0.001; OR = 1.37), ‘Asian’ , ‘glioma’ , ‘PB’ , and ‘HB’ . Supplementary Figures S8–S10 present the forest plot data.For the T/T compared with C/C+C/T model , we alsoP<0.05) except for the subgroups ‘colorectal cancer’ , ‘HB’  under the C/T compared with C/C model; and the subgroup ‘head and neck cancer’  under the carrier T compared with C model. Thus, the T/T genotype of MGMT rs12917 may have been associated with an increased risk of cancer in cases, especially the glioma cases, in the Caucasian population.We did not detect positive results for the other genetic models . A Begg’s funnel plot with pseudo–95% confidence limits under the T/T compared with C/C model is shown in Begg’s and Egger’s tests indicated that results were free of possible publication bias , genotype data were very limited in many subgroup analyses. For instance, we used only three case–control studies in our analyses of the subgroups for gastric ,44,47, bMGMT rs12917 polymorphism in combination with other variants, such as rs2308321 of MGMT, rs25487 of X-ray cross-complementing group 1 (XRCC1), and rs13181 of xeroderma pigmentosum complementation group D (XPD), in certain specific cancers.We did not investigate the genetic effects of the We extracted certain demographic information such as the mean age at diagnosis and the sex of subject, but not other confounding factors such as lifestyle and clinical features. Moreover, we did not perform the relevant stratified meta-analyses due to lack of sufficient usable data.We detected significant heterogeneity amongst studies under the allele T compared with C, C/T compared with C/C, C/T+T/T compared with C/C, and carrier T compared with C genetic models. Complicating factors such as race and cancer type may be sources of inter-study heterogeneity. For instance, we detected decreased levels of heterogeneity in the ‘Caucasian’ and ‘esophageal cancer’ subgroups. Although we observed a positive conclusion in the ‘glioma’ subgroup, we failed to detect reduced inter-study heterogeneity. Only five case–control studies ,30,42,48There may be other undetected or unpublished articles containing potential eligible case–controls in other geographical locations or languages; in other words, our study may suffer from selection bias.MGMT rs12917 and the risk of cancer in general for the T/T compared with C/C and T/T compared with C/C+C/T models. Considering the distinct etiopathogenesis or pathogenesis of different kinds of cancers, more studies of large-scale populations of different ethnicities are required for a more scientific elucidation of MGMT rs12917’s functional role in each particular cancer type.Last but most important, our meta-analysis found a positive conclusion between Some of the limitations to our meta-analysis are as follows:MGMT rs12917 polymorphism is likely to be associated with an enhanced susceptibility to cancer overall, especially glioma, in the Caucasian population.To sum up, our updated pooling analysis offered additional evidence that


The intercellular adhesion molecule‐1 (ICAM‐1)/leukocyte function associated antigen‐1 (LFA‐1) adhesion system regulates leukocyte interactions, migration, and adhesion, and appears to play an important role in atherosclerosis and thrombosis. Therefore, single nucleotide polymorphisms (SNPs) of the ICAM‐1 gene may strongly influence the expression and biological activity of ICAM‐1 and play a potentially important role in the pathogenesis of ischemic stroke. In the current meta‐analysis, we investigated the relationship between the ICAM‐1 gene K469E SNP and the risk of ischemic stroke.Two investigators independently searched PubMed, Web of Science, Google Scholar, WANFANG, China National Knowledge Infrastructure (CNKI) and J‐STAGE for studies published from January 2000 to February 2019 without language restriction. The association of K469E polymorphism and ischemic stroke in three genetic models  were evaluated using Pooled odds ratios (ORs) with 95% confidence intervals (CIs).p = 0.51; I2 = 93%) or recessive model  or dominant model . However, in all three genetic models, subgroup analysis revealed that the K469E polymorphism of the ICAM‐1 gene is associated with ischemic stroke in the Caucasian population.Our study included 20 studies from four continents and four different countries, including 3,137 cases and 15,382 controls. Meta‐analysis results did not show a significant association between K469E polymorphism of ICAM‐1 gene and ischemic stroke when assuming allelic model (OR: 1.12; 95% CI: 0.8 to 1.55; K469E polymorphism of ICAM‐1 gene might be a risk factor for ischemic stroke in Caucasians, which suggested that K469E polymorphism might help in early identification of those at risk and help in primary prevention of ischemic stroke. Approximately, 80% of new strokes are ischemic in origin, 51% stroke death, and 58% of stroke Disability‐Adjusted Life Year (DALY) are because of ischemic stroke  statement , and J‐STAGE for studies published January 2000–February 2019. Searches were conducted using the keywords ‘“ICAM‐1 gene polymorphism”’ or “K469E polymorphism “or “rs5498 polymorphism” or ‘“Lys469glu polymorphism”’ in combination with “stroke” or ‘“ischemic stroke”’ or ‘“cerebrovascular disease”’ or ‘“cerebral infarction”’. Titles and abstracts were then screened and the full text of all the papers that were considered to possibly meet the inclusion criteria was obtained. The cited references of retrieved articles were manually checked to identify any additional eligible studies. Two authors (GN and YK) screened and retrieved reports and excluded irrelevant studies. Any uncertainty about the eligibility of these studies were solved by discussion with all the authors.2.3We included studies that were conducted on human subjects and the studies were searched without any limitations on language. In this meta‐analysis, all included studies met the following criteria: (a) case–control studies focused on the association between K469E polymorphism and ischemic stroke susceptibility; (b) patients with ischemic stroke were diagnosed with neuroimaging confirmed by neurologist; and (c) there were sufficient data of the genotypes in the case–control groups to evaluate the ORs and 95% CIs. The exclusion criteria were: (a) publications with overlapping cases and controls from the similar study; (b) no genotypic data available; and (c) publications with patients with cardioembolic ischemic stroke or hemorrhagic stroke.2.4The relevant data from each study were independently extracted by two reviewers (GN and YK) using a standardized, structured form including the first author, type of design, site of study, patient ethnicity, year of publication, sample size of cases and controls, control design, Hardy–Weinberg equilibrium, source of DNA, genotyping method, and data of the genotypes in the case–control groups. The methodological quality of each study was assessed independently by two reviewers (GN and YK) using the parameters previously used by , the dominant model (EE + EK vs. KK), and recessive model (EE vs. EK + KK). Subgroup analyses were conducted according to ethnicity and methodological quality evaluation. Sensitivity analysis was performed to examine the stability of analysis. Meta‐regression was performed to determine whether methodological quality of study have role in effect size associated with K469E polymorphism. Egger's linear regression test and Begg's test were used for the identification of publication biases. Meta‐analysis, Meta‐regression, Egger's linear regression test, and Begg's test were performed in Comprehensive Meta‐Analysis software .The association between K469E polymorphism and ischemic stroke susceptibility was evaluated by calculating the pooled ORs and 95% CIs. Heterogeneity between the included studies was first determined with Cochran's 33.1The results of the systematic literature search and selection are summarized in Figure 3.2Our research included 20 studies from four continents and four different countries and included 3,137 cases and 15,382 controls. Fifteen studies were from China Geng, , two stu3.3p = 0.51; I2 = 93%) or recessive model  or dominant model . Forest plot of the result for allelic, recessive, and dominant model is demonstrated in Figures p = 0.24; Begg's test: p = 0.2), recessive model , and dominant model . In addition, on inspection of funnel plot, publication bias was not obvious. Funnel plot for detection of publication bias in allelic, recessive model, and dominant model is demonstrated in Figures Due to high heterogeneity within studies in all three genetic models, we used random‑effects model to estimate pooled ORs and 95% CIs. Meta‐analysis results did not show a significant association between K469E polymorphism of ICAM‐1 gene and ischemic stroke when assuming allelic model . Meta‐regression plot is shown in Figures For all three genetic models, we sequentially excluded single study from the pooled analysis and recalculated the summary ORs to check whether the summary ORs were significantly changed. The recalculated ORs were similar indicating stability of analysis. We performed subgroup analysis based on ethnicity and source of recruitment of controls. In all three genetic models, subgroup analysis showed that K469E polymorphism of ICAM‐1 gene was associated with ischemic stroke in the Caucasians population. Similarly in all three models of inheritance, the subgroup analysis based on source of recruitment of controls showed that K469E polymorphism was associated with ischemic stroke in subgroup of studies that recruited controls from hospital setting. The details of subgroup analysis can be seen in Table 4Atherosclerosis is a consequence of chronic inflammation and healing response of the arterial wall to endothelial injury Libby, . DysfuncThe K469E polymorphism is a non‐synonymous SNP located in the fifth immunoglobulin‐like domain of ICAM‐1, which is required for protein dimerization, surface presentation, and solubilization . Significant heterogeneity was also found in all subgroups in all genotype models (I2 ranged from 53% to 94%), except in hospital based subgroup of dominant model. There could be several reasons for these heterogeneities. First, most of these studies had small sample size. Second, different strategies were applied in various studies, including control source, population stratification, phenotype selection, genotyping methods, etc. Moreover, heterogeneity may also be due to the variations in ethnicity, age, and environmental factors.Statistically significant heterogeneity across studies was observed in all genotype models (Our study has several advantages. First, this is the largest meta‐analysis on this issue, including 20 studies and 3,137 cases and 15,382 controls. Second, we included all Caucasian, black, and Asian populations to demonstrate the association between the K469E polymorphism and ischemic stroke. Third, we conducted a comprehensive search, not limited to English literature. We searched the Chinese database and extracted 13 papers in Chinese. Fourth, in most of the included studies, control subjects did not deviate from the Hardy–Weinberg equilibrium. Fifth, we performed our analysis in allelic, recessive, and dominant model of inheritance. Finally, in our analysis, publication bias is not obvious.There were a few limitations in our study. First, some studies included in the meta‐analysis had small sample size and may have provided inconsistent results due to low statistical power. Second, stroke risk varies as per specific subtypes of stroke; however, none of the studies included in the meta‐analysis presented the data of subtype of stroke; therefore, meta‐analysis based on stroke subtype could not be performed. Third, significant heterogeneity was evident in our meta‐analysis. Lastly, multiple factors can influence the development of ischemic stroke, especially the gene‐gene interaction and gene‐environment interaction, which were not analyzed due to lack of sufficient data.Despite these limitations, our meta‐analysis indicated that Caucasians with the ICAM‐1 gene K469E polymorphism are susceptible to ischemic stroke. In order to explore clear conclusions, further elaborate large‐scale epidemiological studies, including all ethnic groups from different geographic locations are required to validate these findings.5Evidence suggests that K469E polymorphism of ICAM‐1 is a risk factor for ischemic stroke susceptibility among Caucasians. There is hope that K469E polymorphism may serve as a marker of risk for multigenic complex disease ischemic stroke and might help in early identification of at risk individuals and form preventive strategies. Considering the importance of ICAM‐1 pathogenesis of ischemic stroke, further larger studies with gene–environment interactions in diverse ethnicity should be conducted to clarify the association between K469E polymorphism and ischemic stroke susceptibility. Furthermore, the underlying molecular causal pathways that confer susceptibility to ischemic stroke are warranted to be established to identify target for prevention and treatment of ischemic stroke.The authors declare no conflicts of interest.GN and YK designed the study. GN, YK and JY carried out literature search, review and selection. GN and YK carried out the statistical analysis and drafted the manuscript. All authors were involved in revising the manuscript critically for important intellectual content. All authors read and approve the final manuscript.This article does not contain any studies with human participants or animals performed by any of the authors.For this type of study formal consent is not required.


Background: Several studies investigated the relationship between antisense non-coding RNA in the INK4 locus (ANRIL) variants and the risk of ischemic stroke (IS), yet whether ANRIL variants are associated with IS remain controversial. Therefore, we performed the present study to obtain a more conclusive result. Methods: Literature retrieval was conducted in PubMed, Medline and Embase. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated. Results: Eighteen studies were enrolled for analyses. Pooled overall analyses showed that rs2383206  and rs10757274  variants were significantly associated with an increased risk of IS. Further subgroup analyses by ethnicity revealed that rs2383206, rs10757274 and rs10757278 variants were all significantly correlated with an increased risk of IS in Asians. Additionally, rs10757278 polymorphism was also significantly correlated with an increased risk of IS in Caucasians. Conclusions: Our findings indicated that rs2383206, rs10757274 and rs10757278 variants may impact individual susceptibility to IS in Asians. Moreover, rs10757278 polymorphism may also impact individual susceptibility to IS in Caucasians. Ischemic stroke (IS) is one of the leading causes of morbidity and mortality all over the world . So far,ANRIL) is located on human chromosome 9p21, a region that has been repeatedly linked to atherosclerosis and its associated ischemic vascular diseases [CDKN2A), CDKN2B and methylthioadenosine phosphorylase (MTAP) are modulated by ANRIL. It was shown that the above-mentioned proteins were abundantly expressed in atherosclerotic lesions, and they could promote atherosclerosis by impacting vascular remodeling, thrombogenesis and plaque stability [ANRIL could promote inflammation by inhibiting caspase recruitment domain family member (CARD) 8 and activating the NF-κB pathway [ANRIL in regulating atherosclerosis and inflammation as well as the close relationship between these two processes and IS, it is believed that functional ANRIL variants may also be involved in the development of IS.Antisense non-coding RNA in the INK4 locus  guideline . The autANRIL variants and IS; (b) it provides genotypic and/or allelic frequency of ANRIL variants in cases and controls; (c) full text in English or Chinese available. Studies were excluded if one of the following criteria was fulfilled: (a) not related to ANRIL variants and IS; (b) case reports or case series; (c) abstracts, reviews, comments, letters and conference presentations. For duplicate reports, we only included the study with the largest sample size for analyses.To test the research hypothesis of this meta-analysis, included studies must meet all the following criteria: (a) case–control study on correlation between ANRIL variants in cases and controls. Additionally, the probability value  of Hardy–Weinberg equilibrium (HWE) was also calculated. When necessary, we wrote to the corresponding authors for raw data. We used the Newcastle–Ottawa scale (NOS) to evaluate the quality of eligible studies [The following data were extracted from included studies: (i) name of the first author; (ii) publication year; (iii) country and ethnicity; (iv) sample size; and (v) genotypic distribution of  studies . This scP-values ≤0.05 were considered to be statistically significant. Between-study heterogeneities were evaluated with 2I statistic. Random-effect models (REMs) would be used to pool the data if 2I ≥ 50%. Otherwise, fixed-effect models (FEMs) would be employed for synthetic analyses. Subgroup analyses by ethnicity were subsequently performed. Sensitivity analyses were executed to test the stability of synthetic results. Funnel plots were used to assess publication biases.All statistical analyses were achieved using Review Manager Version 5.3.3 . Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to estimate strength of associations, and We found 115 potential relative articles. Among these articles, a total of 18 eligible studies which met our inclusion criteria were included for synthetic analyses see . The NOSANRIL variants and the risk of IS, six studies about rs1333040 polymorphism (2552 cases and 3359 controls), four studies about rs1333049 polymorphism (1543 cases and 2888 controls), five studies about rs2383206 polymorphism (2987 cases and 3600 controls), six studies about rs2383207 polymorphism (3793 cases and 8372 controls), seven studies about rs10757274 polymorphism (3106 cases and 16440 controls) and ten studies about rs10757278 polymorphism (6051 cases and 24199 controls) were enrolled for analyses. Significant associations with the risk of IS were detected for rs2383206  and rs10757274  variants in overall analyses. Further subgroup analyses by ethnicity of participants revealed that rs2383206, rs10757274 and rs10757278 variants were all significantly correlated with the risk of IS in Asians. Moreover, rs10757278 polymorphism was also significantly correlated with the risk of IS in Caucasians .ANRIL variants and the risk of IS. Our overall and subgroup analyses suggested that rs2383206, rs10757274 and rs10757278 variants were all significantly associated with an increased risk of IS in Asians. In addition, rs10757278 polymorphism was also significantly associated with an increased risk of IS in Caucasians. As shown in Supplementary Figure S1, for rs1333040, rs1333049, rs2383206 and rs2383207 variants, between-study heterogeneities were trivial, and thus pooled analyses were mainly performed with FEM. For rs10757274 and rs10757278 variants, however, obvious between-study heterogeneities were observed for recessive and additive comparisons, and thus REMs were employed for these analyses.To the best of our knowledge, this is the most comprehensive meta-analysis on associations between ANRIL variants and IS may be ethnic-specific, and this may explain why inconsistent results were observed in included original studies, especially when these studies were performed in different populations.There are several points that need to be addressed about this meta-analysis. First, the exact function of ANRIL is still unclear, and therefore the underlying mechanisms of our positive findings need to be investigated by future investigations. Second, the pathogenic mechanism of IS is rather complex, and it is unlikely that a single genetic variant can significantly contribute to its development. So to better illustrate potential correlations of certain genetic variants with IS, we strongly recommend further studies to perform haplotype analyses and explore potential gene–gene interactions. Third, it is also worth noting that according to our findings, the associations between ANRIL variants and IS may also be modified by gene–environment interactions. However, most eligible studies ignore these potential interactions, which impeded us to perform relevant analyses accordingly [As with all meta-analysis, the present study certainly has some limitations. First, our results were derived from unadjusted analyses, and lack of further adjusted analyses for age, gender, smoking status and co-morbidity conditions  may impact the reliability of our findings since the above-mentioned variables may also impact the individual susceptibility to IS ,32. Secoordingly ,36. On aIn conclusion, our meta-analysis suggested that rs2383206, rs10757274 and rs10757278 variants may impact individual susceptibility to IS in Asians. Moreover, rs10757278 polymorphism may also impact individual susceptibility to IS in Caucasians. However, considering that the sample sizes of several comparisons were still relatively small, further well-designed studies with larger sample sizes are still warranted to confirm our findings.