Several systematic reviews and meta-analyses demonstrated the association between depression and the risk of coronary heart disease (CHD), but the previous reviews had some limitations. Moreover, a number of additional studies have been published since the publication of these reviews. We conducted an updated meta-analysis of prospective studies to assess the association between depression and the risk of CHD.Relevant prospective studies investigating the association between depression and CHD were retrieved from the PubMed, Embase, Web of Science search (up to April 2014) and from reviewing reference lists of obtained articles. Either a random-effects model or fixed-effects model was used to compute the pooled risk estimates when appropriate.Thirty prospective cohort studies with 40 independent reports met the inclusion criteria. These groups included 893,850 participants  during a follow-up duration ranging from 2 to 37 years. The pooled relative risks (RRs) were 1.30  for CHD and 1.30  for myocardial infarction (MI). In the subgroup analysis by follow-up duration, the RR of CHD was 1.36  for less than 15 years follow-up, and 1.09  for equal to or more than 15 years follow-up. Potential publication bias may exist, but correction for this bias using trim-and-fill method did not alter the combined risk estimate substantially.The results of our meta-analysis suggest that depression is independently associated with a significantly increased risk of CHD and MI, which may have implications for CHD etiological research and psychological medicine.The online version of this article (doi:10.1186/s12888-014-0371-z) contains supplementary material, which is available to authorized users. Cardiovascular disease (CVD) is now a public health crisis in both developed and developing countries. It is the leading cause of mortality and extorts heavy social and economic costs globally ,2. AlthoDepression is a common mental disorder, which usually causes severe disability and imposes a huge burden of disease on individuals, families and societies . It is ePrevious reviews -14 showeWe performed this systematic review in accordance with the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines . A literStudies were included in the meta-analysis if they met the following criteria: (1) the study was a population-based or community-based prospective cohort study design; (2) the exposure of interest was depression symptoms; (3) the outcome of interest was CHD or MI; (4) participants were free of CHD at study entry; (5) the study reported the risk estimates with 95% confidence interval (CI) for the association between depression and CHD. Studies were excluded if (1) the study had a retrospective design; (2) the estimates were presented without standard errors or other information that allowed calculation of standard errors; (3) no confounders were adjusted for. In the case of multiple publications from the same study, only the most recent paper or article with a longer follow-up was included.We extracted the following information from each retrieved study: name of the first author, year of publication, study location, characteristics of study population at baseline, duration of follow-up, definition and measurement of exposure, outcomes, number of cases, size of cohort, adjusted relative risk (RR) with 95% CI, and covariates that were adjusted in the multivariable analysis. Data extraction was conducted independently by two authors (Y.G. and Y.H.G). Interobserver agreement was assessed using Cohen kappa (κ), and any differences were resolved by discussion with the third author (Z.X.L.).To determine the validity of included studies, two reviewers (Y.G. and Y.H.G.) independently performed the quality assessment by using the Newcastle-Ottawa Scale , which iIn this meta-analysis, the RR was used as a common measure of the association between depression and the risk of CHD, and the hazard ratios (HRs) was considered equivalent to RRs. If articles provided RRs for women and men separately, we pooled both risk estimates to obtain one overall estimates for the primary analysis. If studies reported fatal or nonfatal outcomes, the two outcomes were considered as two independent reports respectively for all analysis. To be consistent across studies, we used binary variables (yes/no) for depression and CHD. We did not include studies using depressive scale as a continuous variable because the risk estimates were not comparable with those using categorized depression measures.Q test and I2 statistic. P < 0.10 was considered indicative of statistically significant heterogeneity. Statistical heterogeneity among studies was evaluated using the I2 statistic, where values of 25%, 50% and 75% represent cut-off points for low, moderate and high degrees of heterogeneity, respectively [The heterogeneity among studies was estimated by the Cochran ectively . When apectively .We conducted stratified analyses and sensitive analyses to evaluate the influences of study and population characteristic on study results. We used the Begg test , the EggInitially, 9022 articles from the PubMed, Embase, and Web of Science were identified. The majority were excluded after the first screening of titles or abstracts, because they were duplicates, reviews, case–control studies, cross-sectional studies, or not relevant to our analysis. After assessing full texts for detailed evaluation, 30 studies -52 met tAdditional file .Thirty studies with 39 reports were included in the analysis of depression and CHD risk. The results from random-effects meta-analysis of depression and the risk of CHD were shown in Figure P = 0.001; I2 = 64%). Visual inspection of the Begg funnel plot did not identify substantial asymmetry. Neither the Begg test nor the Egger test for publication bias reached significance (P > 0.05 for both tests).The results from random-effects meta-analysis of depression and the risk of MI were shown in Figure Of note, one study  was exclTable P =0.001, I2 = 52.8%). Exclusion of the study by Cohen et al. that enrolled patients with hypertension yielded a pooled RR of 1.29 . We excluded any single study in turn and pooled the results of the remaining studies. The overall combined RR did not change substantially, with a range from 1.29  to 1.34 .Sensitivity analyses were performed to detect potential sources of heterogeneity in the association between depression and the risk of CHD, and to examine the influence of various exclusion criteria on the combined risk estimate. Exclusion of 3 studies that their outcomes were CHD plus MI showed a greater risk 1.35 , and statistical heterogeneity was forcefully attenuated , which identified no evidence of substantial publication bias, but the Egger test did not (P = 0.059). A sensitivity analysis using the trim-and-fill method was performed with 5 imputed studies, which produced a symmetrical funnel plot . Correction for potential publication bias therefore did not alter the combined risk estimate substantially.Visual inspection of the Begg funnel plot did not reveal any asymmetry. The Begg test was not significant , 1.30  for CHD and MI, respectively, which were much lower than those from a previous meta-analysis published 2007 for CHD  and for MI . Our meta-analysis included 19 new prospective cohort studies with larger sample size and many more cases, which significantly enhanced statistical power to detect the potential associations of depression with CHD and MI. In addition, we also fully explored the potential publication bias. Although potential publication bias may exist, correction for this bias using trim-and-fill method remained statistically significant. More important, compared with the previous meta-analysis in 2007, the associations differed between populations of different ethnic backgrounds was investigated. We found that the association trended to be stronger for participants from the United States than for European participants. There was null statistically significant risk of CHD in Hong Kong, which might result from the limited number of included studies . Given that the studies included in our meta-analysis were conducted in affluent countries (areas) in North America, Western Europe, and Hong Kong, the results should not be extended to developing countries. In order to make the finding generalize to other populations, more studies conducted in other populations from Asia, Africa and South America are warranted.Our subgroups analyses identified two important and valuable findings. A major finding was that depression increased the risk of CHD in the group of less than 15 years follow-up, but did not have statistically significant association for equal to or more than 15 years follow-up, which was a very interesting phenomenon. One possible explanation to this finding was that most studies have usually measured depression only one time at the beginning of the study, and have assessed the outcomes at the end of the follow-up. It was well known that depression was treatable and depressive individuals could recover during the follow-up. The longer the follow-up duration, the more people could recover, which might weaken the association between depression and CHD risk. In turn, the finding confirmed that the depression was associated with CHD risk.Another finding was that depression could be an independent risk factor for CHD. In our subgroup analyses, studies adjusted for smoking, BMI, hypertension, diabetes, physical activity, and socioeconomic status did not influence the result of positive association, which suggested that adequate adjustment was not definite to weaken the pooled effect estimate and conclusion .In the subgroup analysis type of depression measurement, it was notable that the association was much stronger in these studies that identified depression using self-reported scales rather than structured clinical diagnostic interviews or clinical diagnosis. One possible interpretation was that the estimates of depression may differ depending on the use of dimensionally versus categorically based depression assessment tools . StructuOn the basis of our research findings, it was expected that treatment with antidepressants would reduce the risk of development of CHD. Conversely, some observational studies ,61 suggeInterestingly, although the evidence from observational studies on the association between depression and the risk of CHD seemed to be robust, the evidence from randomized controlled trials (RCTs) did not find an effect of the intervention ,65. We nOur review is very valuable and crucial though it is an updated meta-analysis. Firstly, we not only included the prospective cohort studies, also included 19 more new studies and 759,922 more new participants than the previous reviews, which provided stronger and more sufficient evidence. Secondly, on the basis of our subgroup analysis, an important methodological finding was worth paying attention to. In cohort studies with longer follow-up, exposure (such as depression) might be mutative, but the exposure measurement frequency was often inadequate, which should arouse the investigator’s attention in the population-based observational epidemiological studies.A few limitations of our meta-analysis should be acknowledged. Firstly, we observed robust and consistent associations across different subgroups via sensitivity analyses and subgroup analyses. Yet as a limitation, there was the evidence of heterogeneity across the studies used for the analysis of association between depression and the risk of CHD. The heterogeneity might result from the difference of participants’ characteristic, sample sizes, study designs, and diagnostic criteria of depression. Thus, the results of this meta-analysis should be interpreted cautiously. Secondly, the measurement of depression mostly used the self-reported symptom scales, which could cause the misclassification of exposure , and migOn the basis of our findings, we put forward some suggestions for future studies. Firstly, more studies are needed to assess depression in a repeated-measures design with diagnostic interview as well as self-report during the follow-up. Secondly, in order to achieve stronger research evidence, investigators should strive to improve the standardization of their depression measurement and outcome definitions . ThirdlyIn conclusion, our meta-analysis of prospective studies suggests that depression is probably an independent risk factor, which is associated with a 30% increase risk for CHD and MI. Given the high prevalence of depression in the general population and the heavy economic burden of CHD, it is greatly important for us to take depression account into the clinical prevention and treatment of CHD.


Psoriasis is considered a systemic inflammatory disorder. Previous studies have reported conflicting positive or negative correlations between psoriasis and chronic obstructive pulmonary disease. We performed a meta-analysis to determine whether there is an associated risk between psoriasis and chronic obstructive pulmonary disease. We performed a complete 30-year literature search of MEDLINE, Embase, and Cochrane Central Register databases on this topic. Four observational studies with a total of 13,418 subjects were identified. The odds ratios of chronic obstructive pulmonary disease in subjects with psoriasis/mild-to-moderate psoriasis were analyzed using the random-effects model, while the odds ratios of chronic obstructive pulmonary disease in subjects with severe psoriasis and current smoking in subjects with psoriasis were analyzed using the fixed-effect model. We found that psoriasis patients were at a greater risk of developing chronic obstructive pulmonary disease than the general population  and that the association between of psoriasis and with chronic obstructive pulmonary disease was stronger among patients with severe psoriasis . Psoriasis patients should be advised to cease smoking to reduce their risk of COPD. Moreover, identification of this potential risk may enable earlier implementation of preventive measures for reduction comorbidity and mortality rates. Psoriasis is a common chronic and relapsing immune-mediated inflammatory disease of the skin that affects approximately 2–4% of the population worldwide. The cliChronic obstructive pulmonary disease (COPD), which encompasses chronic obstructive bronchitis and emphysema, affects approximately 10% of the general population. A progrIt is widely accepted that common pathogenic mechanisms are shared among many human chronic inflammatory diseases of unrelated pathology and manifestation. Increasing our understanding of the strength of the correlation between psoriasis and COPD will help ensure that future observational studies include adequate adjustments for the presence of COPD among patients with psoriasis. The purpose of this review was to examine the association between psoriasis and COPD using a meta-analysis.The review protocol was registered in the PROSPERO database before the start of the review process (CRD42015025224).To identify relevant psoriasis studies that included COPD as an outcome measure, three reviewers  systematically searched the MEDLINE, Embase, and Cochrane Central Register databases using the search terms psoriasis, COPD, and chronic obstructive pulmonary disease. Papers published in English and dated between January 1980 and December 2014were included in this study.To determine eligibility for inclusion in this review, we screened abstracts using the criteria of case-control, cross-sectional, cohort, or nested case-control design studies examining COPD in relation to psoriasis with no limits on participant age, sex, or nationality. The selection criteria for inclusion were as follows: (i) human-only studies; (ii) provision of original data; (iii) inclusion of a reference group; (iv) provision of odds ratios (ORs), risk ratios, or hazard ratio estimates with confidence intervals (CIs) ; or consideration of COPD as a specific outcome event.In this study, we identified 43 articles from the initial search  and throThree reviewers independently collected the following descriptive data for each included study: (i) first author; (ii) study characteristics ; (iii) participant characteristics ; and (iv) outcome characteristics (diagnostic criteria of psoriasis).The Newcastle-Ottawa Scale was used2 tests. An I2 value > 50% was considered as substantial heterogeneity, and DerSimonian and Laird random-effect models were used to compute the globalOR. However, if the inter-study heterogeneity was not substantial (I2< 50%), a fixed-effect model was considered suitable. To investigate possible reasons for heterogeneity, we performed subgroup analysis and meta-regression using pre-specified variables and random-effects meta-analysis. Pre-specified sources of heterogeneity in meta-regression included study location, source population, study design, study quality, severity of psoriasis, psoriatic arthritis included and outcome ascertainment. The methods and findings of this review strictly followed the Meta-analysis of Observational Studies in Epidemiology group guidelines and checklist[http://ims.cochrane.org/revman). Meta-regression was performed using STATA version 10.0 .The primary outcome was the association of psoriasis with the risk of COPD for each study. The degree of heterogeneity between studies was assessed using Ichecklist. The CocOf a total of 44 studies, four observational studies conducted in Europe (one study), Middle East (two studies), and East Asia (one study) in inpatient and outpatient settings, were selected through a systematic review and included in this meta-analysis–14. The 2 = 96%). However, low heterogeneity within each separate subgroup  were displayed. A random-effects modeling analysis of the pooled data from the four studies showed a significant effect between psoriasis patients and controls on the associated prevalence of COPD   . An opti36–2.65) .2 = 62%; severe psoriasis, I2 = 0%), and a significant effect of the associated prevalence of COPD between severe patients and controls , while the pooled data showed a significant effect between psoriasis patients and controls in the prevalence of current smoking   . RegrettOur results demonstrated that psoriasis patients are at a greater risk of developing COPD , and the association was stronger in patients with severe psoriasis .This study collected and reanalyzed individual participant data including a total of 13,418 participants from four observational studies for the effects of psoriasis on the incidence of COPD. These studies provide almost all of the available epidemiological evidence worldwide on the topic. Since the 1980s, only five studies–14, 17 hThe consistent association between psoriasis and COPD suggests a likely pathophysiologic link between the two diseases. The hypothesis of a common cytokine-based pathology, wherein one inflammatory autoimmune disease significantly increased the risk of another, is gradually being accepted. It is bIt is interesting that the pooled OR of the associated prevalence of COPD between the patients with severe psoriasis and controls was significant compared to those with mild-to-moderate psoriasis. Likewise, psoriasis confers an increased risk of developing cardiovascular and ischemic heart disease; only severe psoriasis has been reported as an independent cardiovascular risk factor. UnsurprThis study has several strengths. We performed a careful meta-analysis using three different data sources. Multiple authors reviewed the data independently, thereby minimizing the risk of publication bias or missing data. By separating prevalence in the analysis, as well as the severity of psoriasis, we were able to better quantify the possible temporal association between psoriasis and COPD. We also performed meta-regression analyses to identify sources of heterogeneity that commonly occur during meta-analyses of observational data. While conducting this study, we were fully aware of several well-documented limitations, including the distorting effects that publication and location bias may have on systematic reviews and meta-analyses. Only foMeanwhile, COPD is more prevalent in patients of middle and older age: the mean age of participants from the three included studies was 44.3 years, which almost coincides with the COPD onset age–14. HoweIn summary, patients with psoriasis are at greater risk of developing COPD. The association between psoriasis and COPD was stronger in patients with severe psoriasis. Physicians should be aware of this potential risk to reduce comorbidity and mortality rates. Evidence of the link between psoriasis and COPD is strengthened by the fact that a number of monoclonal antibodies and/or biological reagents currently under clinical development have been designed for use in patients with psoriasis or COPD.S1 MOOSE Checklist(DOC)Click here for additional data file.


Although the polymorphisms of PTPN22 and the variants of CTLA-4 have been reported to be the susceptibility genes, which increased risk of latent autoimmune diabetes in adults (LADA), the results remained inconclusive. The aim of this meta-analysis was to evaluate the association between the polymorphisms of two genes and LADA. We performed a systematic review by identifying relevant studies and applied meta-analysis to pool gene effects. Data from ten studies published between 2001 and 2013 were pooled for two polymorphisms: rs2476601 in the PTPN22 gene and rs231775 in the CTLA-4 gene. Data extraction and assessments for risk of bias were independently performed by two reviewers. Fixed-effect model and random-effect model were used to pool the odds ratios; meanwhile, heterogeneity test, publication bias and sensitive analysis were explored. The minor T allele at rs2476601 and the minor G at rs231775 carried estimated relative risks (odds ratio) of 1.52 (95 % CI 1.29–1.79) and 1.39 (95 % CI 1.11–1.74), respectively. These alleles contributed to an absolute lowering of the risk of all LADA by 4.88 and 14.93 % when individuals do not carry these alleles. The estimated lambdas were 0.49 and 0.63, suggesting a codominant model of effects was most likely for two genes. In summary, our systematic review has demonstrated that PTPN22 rs2476601 and CTLA-4 rs231775 are potential risk factors for LADA. An updated meta-analysis is required when more studies are published to increase the power of these polymorphisms and LADA.The online version of this article (doi:10.1007/s00592-014-0613-z) contains supplementary material, which is available to authorized users. Latent autoimmune diabetes in adults (LADA) is commonly considered as a type of autoimmune diabetes that resembles type 1 diabetes (T1D); however, it masqueraded as type 2 diabetes (T2D) in the initial stage . It is cThe patients with LADA were present autoimmunity, immune-mediated β-cell dysfunction and damage as part of their disease process. The progression to insulin dependence in LADA patients is believed more rapidly than classic type 2 diabetes patients who were negative for islet autoantibodies that have been proved with no progressive damage in beta cell . HoweverIt has been clearly identified that there is a strong genetic component affects diabetes. Genome-wide association studies (GWAS) have had considerable success in identifying genetic contributions to T1D and T2D. Unfortunately, LADA is not arousing our attention, and the genetic studies of LADA are sorely lacking. However, some newly articles reported that the single nucleotide polymorphism (SNP) of some genes that associated with T1D and T2D is also showed relevancy with LADA . The pro, [(PTPN 1858 or rs2476601)], [(CTLA-4 +49A/G or rs231775)] and [(LADA or latent autoimmune diabetes in adults)]. Only published articles were considered and set no restriction on the source of controls. We browsed the title and abstract of all related manuscripts, manually examined reference lists for additional citations and obtained the full text of all potentially relevant articles. If there were more than one articles published by the same content, we choose the most complete and recent study.Two reviewers (F.D. and W. K. L.) independently went through all titles and abstracts of the identified studies. Studies were selected if they met the following criteria: a case–control study that were written in English or Chinese; genotyped PTPN22 (PTPN22 1858 or rs2476601) or CTLA-4 (+49A/G or rs231775) polymorphisms and detailed data of each genotype; the outcome was latent autoimmune diabetes in adults (LADA); articles had to report the odds ratio and corresponding 95 % confidence interval or provided the sufficient information for estimation. Studies with insufficient data for pooling that with no frequencies of genotypes for each polymorphisms and outcomes were excluded.For quality control, information was extracted from the studies independently by two investigators (F.D. and H. W. P.). If lack of genotype information, we will try to contact the corresponding author in order to obtain required data. If they did not provide data, those studies were excluded from our review. General characteristics  of included studies were extracted. Any disagreement was resolved by consensus.The quality of studies was also independently assessed by the same reviewer (F.D. and G. Y.) based on a risk of bias score for genetic association. This was modified on the basis of both traditional epidemiologic considerations and genetic issues, which were developed by Thakkinstian et al. . The scoP value less than 0.05, it was considered to be disequilibrium. We performed both per-allele and per-genotype approaches to estimate the strength of association between the polymorphism of genes and LADA risks.We used the Comprehensive Meta-Analysis software (ver2.0) for all statistical analyses. The Hardy–Weinberg equilibrium (HWE) was examined in control groups by Fisher’s exact test. If the study was found not to be in HWE with P value was used to determine whether the overall gene effect was significant (α = 0.05). Heterogeneity of odds ratios across studies was calculated by a Q test, and the degree of heterogeneity was quantified by I2 test [P > 0.10, a fixed-effect model was selected to pool the data, which can be considered as the evidence of homogeneity between studies. Otherwise, a random-effect model was used. In addition to this, the degree of heterogeneity was quantified using I2  [I2 was greater than 50 % [Suppose that D and d are risk and non-risk alleles, and DD, Dd and dd are minor homozygous, heterozygous and common homozygous genotype, respectively, for each polymorphism. The risk allele frequency in each group was estimated for each study by reported genotype data, and overall prevalence along with 95 % confidence intervals was estimated for each SNP. The Z-test was used to determine the statistical significance of the pooled OR, and its I2 test . If the geneity) . We choohan 50 % . If therhan 50 % , 34. If λ), which is defined as the ratio of logOR2 to logOR1, was then estimated using the model-free Bayesian approach. This parameter ranges from 0 to 1, which represents the heterozygote effect as a proportion of the homozygote variant effect and captures information about the genetic mode of action as follows: If λ = 0, a recessive (DD vs. Dd +dd) model is suggested; if λ = 1, a dominant model (DD + Dd vs. dd) is suggested; and if λ = 0.5, a codominant model  is suggested. If λ > 1 or λ < 0, then a homozygous or heterosis model is likely, although this is rare. The two log odds ratios are modeled as either fixed or random effects, as described in the second statistical analysis enumerated above. Once the best genetic model is identified, this model is used to collapse the three genotypes into two groups and to pool the results again. For lambda, WinBugs 1.4.2 was used with vague prior to distributions for estimation of parameters . The models were run with a burn-in of 1,000 iterations, followed by 10,000 iterations for parameter estimates.We perform the model-free approach to estimate the genotype effect , two oddPublication bias was assessed using the cumulative forest plot and Egger’s regression intercept . CumulatP < 0.001).Thirteen relevant articles were identified after the primary literature search about PTPN22; seven articles were excluded after screening abstracts and full texts. Among these articles, some are described the irrelevant content to the LADA topic and others are lack of the detailed data we required. Finally, only six articles were left. There were 22 studies conform to the standard after the preliminary search about CTLA-4 gene. After extraction, a total of six case–control studies were selected according to the search criteria for LADA related to the polymorphism of CTLA-4. HWE was calculated for control groups in all articles; we found that one study was showed disequilibrium  and 9.9 % (95 % CI 7.6–12.7) in non-LADA group (I2 = 91), which were both estimated by random model. The odds ratios (T vs. C) were not heterogeneous , with a pooled odds ratio of 1.52 (95 % CI 1.29–1.79). The overall gene effect estimated by fixed-effect model was significant (P < 0.001). This suggested that individuals carrying the minor T allele had 52 % increased risk of developing LADA than those carrying the major C allele (Table P = 0.54). The sensitive analysis was performed by omitting one study at a time, which the method was used to make sure that no individual study was entirely responsible for the combined results. From the Table P = 0.00004, Z = 4.12) when α was set to 0.05, which suggest that 21 unpublished negative studies would have to be included to convert the combined P value to a nonsignificant value. The above results show that our results were statistically reliable.There were six case–control studies described the association between PTPN22 rs2476601 polymorphism and LADA, which included 1,088 cases and 4,079 controls , whereas the OR2 for CT versus CC was homogenous . They both calculated by fixed-effect model. The pooled OR1 and OR2 were 1.86 (95 % CI 0.94–3.68) and 1.52 (95 % CI 1.26–1.84), respectively, which suggested that individuals with TT and CT genotypes had 86 and 52 % higher risk of LADA than those carrying CC genotype. The λ = 0.49 (95 % CI 0.07–0.96) which suggested that a codominant effect was most likely, although one genotype effect did not reach statistical significant.Genotype frequency and estimated OR for each study were shown in Table I2 = 95), and in non-LADA group was 44.9 % (95 % CI 36.3–53.9), which estimated by random model (I2 = 97). The pooled OR was calculated by random-effect model due to the high heterogeneity . The odds ratio of G versus A is 1.39 (95 % CI 1.11–1.74) with statistical significance (P = 0.004), which indicated that individuals carrying G allele had 39 % increased risk of developing LADA than those carrying A allele. In cumulative plot, the summary OR was a bit different in the first two studies, whereas not much changed in a smooth curve with the increase of the literature. Egger’s regression test also suggest no publication bias . In the sensitive analysis, after each removed a piece of literature have not seen a big difference in the OR values have changed. Classic fail-safe N value of CTLA-4 rs231775 is 24  when α was set to 0.05, which suggest that 24 unpublished negative studies would have to be included to convert the combined P value to a nonsignificant value. This shows that our results are stable enough. When studies were divided according to the ethnic group, the result showed that there is a significant association in Caucasian. The pooled odds ratio is 1.45 (95 % CI 1.09–1.92), with a significantly statistical gene effect (P = 0.01) but a highly heterogeneity . There is only one Chinese study that belongs to Asian population with no statistical significance .The associations between CTLA-4 rs231775 and LADA were investigated in six case–control studies, with 528 cases and 2,687 controls. The pooled frequency of minor G allele in LADA group was 49.2 % (95 % CI 35.2–63.3), along with high heterogeneity  and OR2 . The summary odds ratios for the GG and AG genotype were estimated by random model, they are 1.96 (95 % CI 1.10–3.50) and 1.68 (95 % CI 1.12–2.53), respectively. These point estimates can be interpreted as that person with the GG and GA genotypes had 96 and 68 % higher risks of developing LADA than persons with the AA genotype. The estimated λ = 0.63 (95 % CI 0.15–0.98) which suggested that a codominant effect was most likely.In heterogeneity test, there is a moderate heterogeneity across OR1  on the LADA. The analyses included pooling data from five and six studies with a total sample size of 4,728 and 3,215 subjects. We were able to identify PTPN22 rs2476601 and CTLA-4 rs231775 polymorphisms as genetic markers that might increase the risk of LADA. Individuals who carried minor allele T in rs2476601 had 52 % increased risk of developing LADA relative to those carrying C allele, while individuals carrying the risk allele G in rs231775 may lead to an increasing risk of having LADA by 39 % compared with allele A. The results suggest association in Caucasians, that is, carriage of G in the CTLA-4 rs231775 increases 45 % relative to carriage of A allele. However, Asian populations showed an unrelated result. This difference may be due to the different genetic backgrounds and limited article.The minor T risk allele of the PTPN22 rs2476601 polymorphism investigated is quite rare in non-LADA group, with frequency of 9.9 %. However, it is high in the LADA group, with frequency of 16.9 %. The PAR for the minor T was 4.88 %, which suggested that PTPN22 rs2476601 polymorphism probably serves as a marker for an absolute lowering of the risk of all LADA in Caucasians by 4.88 % when individuals do not carry T allele. The risk G allele in CTLA-4 rs231775 is common, with similar frequencies of 44.9 and 49.2 % in non-LADA group and LADA group, which might indicate an important effect at a population level. The PAR of CTLA-4 G allele was 14.93 %, which may provide a useful clinical estimation that might contribute an absolute lowering of the risk of all LADA by 14.93 % when individuals do not carry this allele.Genotypic effects were also estimated for PTPN22 rs2476601 and CTLA-4 rs231775. For PTPN22 rs2476601, the estimated OR1 for TT versus CC and OR2 for CT versus CC were 1.86 and 1.52 in Caucasian, respectively, and estimated lambda was 0.49, suggesting a codominant mode of gene effect. However, the 95 % confident interval of lambda laid from 0.07 to 0.96, which suggested that the genetic mode could be recessive dominant and codominant. This pooling was based on small number of included studies, and thus, uncertainty of gene effects was still present. For CTLA-4 rs231775, the genotype effects of GG and GA versus AA were 0.96 and 0.68, respectively. The point estimated lambda was 0.63, suggesting a codominant mode of gene effect. The 95 % confident interval of lambda laid from 0.15 to 0.98, which suggested that the genetic mode could be recessive dominant and codominant.Previous studies have shown that the SNPs of PTPN22 and CTLA-4 are associated with T1DM –43 and ohttp://andis.ludc.med.lu.se). Multiple islet auto-antigens and autoantibodies could be detected before the development of autoimmune diabetes [LADA is a common subgroup of diabetes accounting for about 7 % of all diabetic patients in Europe Below is the link to the electronic supplementary material.


GSTT1 gene encodes a key enzyme involved in the metabolism of xenobiotics and its polymorphisms have been associated with individual susceptibility to various malignancies. Numerous molecular epidemiological studies have been performed to investigate the association between GSTT1 gene polymorphisms and lung cancer susceptibility; however, the results of previous studies were inconsistent. Therefore, the aim of the present study was to conduct a meta-analysis in order to derive a more precise estimation of the association in the East Asian populations. The meta-analysis included 7,415 lung cancer cases and 6,084 controls from 26 published studies in East Asia, which were selected from the PubMed and China National Knowledge Infrastructure databases, up to March 20, 2014. Using crude odds ratios (ORs) with 95% confidence intervals (CIs), a statistically significant association was identified between the GSTT1 null genotype and lung cancer in the East Asian populations . Furthermore, subgroup analyses revealed that the lung cancer risk in smokers carrying the GSTT1 null genotype was significantly increased compared with non-smokers . Thus, the GSTT1 null genotype may increase the risk of lung cancer among the East Asian populations.The  Lung cancer is the most common type of cancer and the leading cause of cancer-associated mortality in the world, becoming a major public health problem globally. Lung cancer accounts for 13% (1.6 million) of the total number of cancer cases and 18% (1.4 million) of mortalities in 2008 (GST θ1 (GSTT1) or GST μ1 (GSTM1) genes (null variants), which result in no enzymatic functional activity  are phase II enzymes that are key in the detoxification of numerous carcinogens . GSTs meactivity ,8. Deficg cancer –12; howeg cancer –15. Therulations ; howeverulations . Thus, tulations ,19–41, aEligible studies were identified by performing a literature search on the PubMed and China National Knowledge Infrastructure (CNKI) databases  using the following keywords: ‘Glutathione S-transferase T1’, ‘GSTT1’, ‘polymorphism’, ‘lung cancer’ and the combined phrases. In addition, the references of all the relevant articles and reviews were searched for additional eligible studies. Furthermore, the full text of each potentially relevant paper was scrutinized to ensure that the following inclusion criteria were met: i) The studies had an observational (case-control or prospective) study design; ii) the authors offered sufficient data for estimating odds ratios (ORs) and their 95% confidence intervals (CIs); and, iii) the patients and controls of all the studies were East Asian . In cases where multiple studies reporting on the same population data met the inclusion criteria, the study with the largest sample size was selected for subsequent analysis.GSTT1 allele were defined as the ‘present’ genotype group, whereas individuals carrying no GSTT1 alleles were classified as the ‘null’ genotype group.In order to minimize bias, the following data were extracted from all the eligible studies by two researchers independently: First author’s surname, year of publication, country, source of the controls , histological type of cancer , smoking status and the different genotypes in the cases and controls. In accordance with the definition used in the majority of previous studies, carriers with at least one The quality of the studies included in the present analysis were assessed using an adapted 10-point Newcastle-Ottawa assessment scale (NOS) . The quaGSTT1 polymorphism was estimated for each study using crude ORs with 95% CIs. The pooled ORs were evaluated for null vs. present genotypes and a χ2-based Q-statistical test was performed to assess the heterogeneity between the studies . However, no statistically significant association was identified in non-smokers . Furthermore, in subgroup analysis by histological type, no statistically significant association was identified for all the stratified analyses. The major results of the meta-analysis and the heterogeneity test are listed in As indicated in As demonstrated in GSTT1 polymorphism was associated with an increased risk of developing lung cancer in the East Asian populations. Furthermore, in subgroups of smoking status, increased lung cancer susceptibility was identified in the smoking population.The results of the current meta-analysis indicated that GSTT1 gene is located on chromosome 22q11.2  compared with Asians (50–60%)  . Thus, eGSTT1 null genotype (GSTT1 gene and tobacco consumption has an effect on the development lung cancer. Furthermore, it has been reported that not all populations were equally susceptible to tobacco-associated carcinogens (GSTT1 null genotype was lower in Caucasians compared with Asians, strengthening the hypothesis that polymorphisms in enzymes that metabolize tobacco carcinogens may have a strong association with ethnicity (Tobacco smoking is known to be one of the major risk factors of lung cancer . Tobaccogenotype ,60,61. Scinogens . The frethnicity .et al (GSTT1 null genotype combined with other GST mutants may have an enhanced risk of developing lung cancer compared with individuals carrying any mutant alone. For instance, specific studies indicated that a daily diet of fruit and vegetables containing isothiocyanates may reduce the risk of lung cancer among carriers of at least one functional GSTT1 allele (et al (GSTT1 alleles. In addition, the results are in accordance with various smaller studies of lung cancer, breast cancer and colorectal adenomas that demonstrated a protective effect of isothiocyanates in GSTT1 null carriers (GSTT1 null individuals.A number of previous studies have evaluated the effects of gene-gene and gene-environment interactions in lung cancer development. Chen et al  indicate1 allele ,64,65. BIn the current meta-analysis, various limitations should be acknowledged. Lung cancer is known to be a consequence of multiple risk factors. For instance, lifestyle, diet, age, gender, environment and ethnicity may all contribute as the possible risk factors . HoweverGSTT1 null genotype may contribute towards the increased lung cancer risk in the East Asian populations. Future large and well-designed epidemiological studies considering the potential interactions are required to confirm the results of the present meta-analysis.In conclusion, the present study indicated that the


BMP4) gene in susceptibility to colorectal cancer (CRC).We performed an updated meta-analysis, using a comprehensive strategy of a logistic regression and a model-free approach, to evaluate more precisely the role of the rs4444235 variant near the Bone morphogenetic protein-4 (G) metric was used to provide a global test of relationship between rs4444235 and CRC risk.A total of 19 studies with 28770 cases and 28234 controls were included. Metagen system with logistic regression was applied to choose the most plausible genetic model for rs4444235. Generalized odds ratio , and pooled per-allele OR was 1.08 (95% CI = 1.05–1.11). Based on the model-free approach, pooled ORG was 1.09 (95% CI = 1.05–1.14) under a random-effect model. Stratified analyses suggested heterogeneity could be in part explained by population ethnicity, study design, sources of controls, and sample size. Sensitivity analysis further supported the robust stability of the current results, by showing similar pooled estimates before and after sequential removal of each study.Metagen analysis suggested the rs4444235 fitted best to an additive model. In assessment of the additive model, heterogeneity was observed (This meta-analysis provides a robust estimate of the positive association between the rs4444235 and CRC risk and further emphasizes the importance of the rs4444235 in CRC risk prediction. Colorectal cancer is a major public health issue in developed countries and is becoming increasingly prevalent in Asia and Africa, with over 1.2 million new cases worldwide each year BMP4), was firstly reported by a meta-analysis of GWAS data to be associated with CRC risk, with a combined OR of 1.11  BMP4 and thus conferred to CRC risk Recently, genome-wide association studies (GWAS) have implicated multiple common single nucleotide polymorphisms (SNPs) in inherited predisposition to CRC Meta-analysis, by integrating published data, may be a powerful tool to clarify the inconsistencies across individual studies. Two meta-analyses have been performed to assess rs4444235 in CRC. The meta-analysis by Li et al. This meta-analysis were conducted according to the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) statement  [14]. GeThe following inclusion criteria should be fulfilled: (1) either case-control or nested case-control studies; (2) clear definition of colorectal cancer cases; (3) studies evaluating relationship between rs4444235 and CRC risk; (4) providing sufficient data to re-calculate the effect metrics, that was, numbers of genotypes in cases and controls. The authors were contacted via E-mail when eligible articles reported insufficient data. If they were unable to provide detailed data, those articles were excluded. Animal studies, reviews, conference abstracts, editorials and letters were excluded. If more than one ethnical population were in one report, each population was considered separately. Studies overlapping with other studies should be excluded, and the one with the most completed information was included. The first study on the association of rs4444235 by Houlston et al. was excluded Data were extracted independently and in duplicate by 2 reviewers (L. Liu & Q. Su). The following data was extracted from each article according to a fixed protocol: the first author, publication year, study design, country, ethnicity, source of controls, numbers of cases and controls, mean age of cases, sex ratio, site/type of colorectal cancer, genotyping method, minor allele frequency (MAF), and frequency of genotypes in cases and controls.2 test (P>0.05). The inverse variance method was applied to estimate the pooled frequency of the risk allele  in various ethnical populations. The genetic effect of the rs4444235 in CRC susceptibility was assessed using the approaches described as below:Hardy-Weinberg equilibrium in controls was re-analyzed using the goodness-of-fit χθ2 and θ3 were estimated using the logistic regression: logit (ijπ) = iα +2zi2θ+3zi3θ, where iα was the indicator of study-specific fixed-effect, ORTC/TT = exp(2θ), and ORCC/TT = exp(3θ). In order to account for an additive component of heterogeneity, a random-effect logistic regression was performed using the GLLAMM module in STATA software via introducing a study-specific random coefficient: logit (ijπ) = iα +(2+νi2θ)i2z+(3+νi3)zi3θ. The most plausible genetic model was determined using the following procedure: if 2θ = 3θ = 0, no significant genetic-association was suggested; if 2θ = 0 and 3θ>0, a recessive genetic model was suggested; if 2θ = 3>θ0, a dominant model was suggested; if 3θ>2>θ0, a co-dominant model was suggested; if 22θ = 3θ, an additive model was likely. In this meta-analysis, the genetic model of rs4444235 was best fitted with an additive model. Then the per-allele OR of the C allele (additive model ) with corresponding 95% confidence interval (95% CI) was estimated in a logistic regression model, by assigning scores of 0, 1, and 2 to the AA, AC and CC genotypes, respectively. Between-study heterogeneity was assessed by the Cochran’s χ2 based Q test and 2I metric. If there was no heterogeneity , a fixed-effect model was used to pool the estimate; otherwise, a random-effect model was applied. To explore the sources of heterogeneity, stratified analyses were performed, if feasible, according to population ethnicity , sources of controls , study design (GWAS and replication study), and total sample size (≤2000 and >2000).Metagen system has provided a general framework to decipher the most plausible genetic model for the rs4444235 that treated the genotypes as independent variables in a logistic regression under both fixed and random effects models G), based on a genetic model-free approach, was also introduced in this meta-analysis G utilized the complete genotype distribution to provide an estimate of overall gene-disease relationship, given that the mutational load was treated as a graded exposure. Heterogeneity was also assessed for ORG metric and stratified analysis was also performed.Additionally, the generalized OR . The pooled frequency of the C allele was 0.463 (95% CI = 0.452–0.474) in Caucasians, similar to that of 0.477 (95% CI = 0.423–0.532) in Asians. Only 1 study was conducted in Africans, and the frequency of the C allele was 0.334.Significant heterogeneity was seen both in Caucasians and Asians. and thus the random-effect model was applied  and 1.18 (95% CI = 1.12–1.25), respectively, suggesting an additive model as the most plausible genetic model. Then the additive model for the rs4444235 was assessed using traditional method. In the additive model, heterogeneity was observed , and thus the random-effect model was applied. The variant was significantly associated with increased CRC risk, with a pooled per-allele OR of 1.08 . Under the random-effect model, significant result was also produced for the association of rs4444235 and CRC risk, with a pooled ORG of 1.09 (95% CI = 1.05–1.14).G assessment , and no significant association was found.When performed stratified analysis by population ethnicity, in Caucasian subgroup of 15 studies, heterogeneity was removed, and the significant association of the rs4444235 still existed for both additive model and ORsessment . HoweverG both showed significant association of rs4444235 with CRC without evidence of heterogeneity, whereas in the hospital-based subgroup of 8 studies, significant heterogeneity was observed and no significant association was reported.According to the sources of controls, in the population-based subgroup of 13 studies, analysis of the additive model and ORG metric, both subgroups showed the positive genetic association with CRC risk, without evidence of heterogeneity. Interestingly, the pooled estimates in the GWAS  were slightly larger than those in the subgroup of replication studies .Regarding to study design, there were 6 GWAS and 13 replication studies. When assessing the additive model and ORG analyses, heterogeneity was removed in the subgroup with large sample size, whereas in the small or moderate size subgroup, heterogeneity still existed. Both subgroups showed the significant association between the rs4444235 and CRC risk.The stratified analysis was also conducted according to total sample size (numbers of both cases and controls), into 2 subgroups: the large sample size subgroup  with 22064 cases and 20876 controls and the small or moderate size subgroup  with 6706 cases and 7358 controls. For both additive model and ORG analysis that no single study significantly altered the pooled ORG. In the Begg’s and the Egger’s tests, there was no evidence of publication bias for both additive model and ORG .Since between-study heterogeneity was observed in this meta-analysis, we further performed sensitivity analysis under the random-effect model. For the additive model, the sensitivity analysis, by sequentially omitting each study, reported a series of pooled OR with 95% CI exceeding 1.00, and the pooled ORs were similar before and after omitting each study . SimilarG metric based on model-free approach Currently, traditional meta-analyses of genetic association studies are usually performed by collapsing genotypes in two categories assuming various genetic models. However, these different models are not independent, and a priori biological justification for the choice of a specific model is seldom available G) analysis showed that CRC cases with higher mutational load than healthy individuals have 9% higher risk for CRC susceptibility. Sensitivity analysis further supported the current results, by showing similar ORs before and after sequentially omitting single study. The positive association of the rs4444235 with CRC risk identified by this meta-analysis was also concordant with the findings of previous meta-analyses In this meta-analysis of 19 case-control studies of 28770 cases and 28234 controls, the metagen analysis indicated that the rs4444235 fitted best to an additive model. Knowledge of the best-fitting model for the rs4444235 may be important in optimizing the use of this SNP in colorectal cancer (CRC) risk prediction. Assessment of additive model indicated that CRC risk was increased by 8% per extra C allele. Based on model-free approach, the generalized OR Click here for additional data file.


TP53 gene in breast cancer risk. The enlightenment of this matter might provide a piece of information about the potential implications of this polymorphism in patient risk. A meta-analysis was conducted considering a large sample size from studies with conflicting results on the R72P polymorphism in breast cancer patients. Relevant studies were selected from PubMed and SciELO databases for data extraction and statistical analysis. Database was built according to the continent and considering the genotype frequencies, sample size and genotyping methodology. The dominant models (RR vs RP + PP and RR + RP vs. PP), homozygous (RR vs. PP), heterozygous (RR vs. RP and RP vs. PP) and the allele (R vs. P) were used. Genotype frequencies were summarized and evaluated by χ2 test of heterogeneity in 2×2 contingency tables with 95% CIs. Odds Ratios (OR) were calculated with a fixed-effect model (Mantel-Haenszel) or a random-effect model (DerSimonian-Laird) if the studies were considered homogeneous (P > 0.05) or heterogeneous (P < 0.05), respectively, using BioEstat® 5.0 software. Supported by a large sample size composed by 25,629 cases and 26,633 controls from 41 studies, we found significant association between the R72P polymorphism in the TP53 gene and the breast cancer risk. The overall data shows an increased risk due to the P allele dominant model, but not in Asia where the risk was associated with the R allele and R dominant model.This study was conducted in order to investigate the implications of the R72P polymorphism in the  The R72P polymorphism in the TP53 gene results of the transversion G → C in the second position of the codon 72 at exon 4. Both the polymorphic alleles vary among ethnic groups   the P variant binds greater to transcriptional machinery  the studies were published from 2002 to 2012; (b) the association between the R72P polymorphism and breast cancer were investigated; (c) the studies were case–control design; (d) genotyping was carried out by molecular biology methods, such as PCR, RFLP-PCR and DNA sequencing; (e) the reference was published in English; (f) histological confirmation of breast cancer diagnosis was performed; and (g) the genotype distributions were available for estimating odds ratios (OR) and 95% confidence intervals (CI).A literature search was conducted in SciELO (Scientific Eletronic Library Online) and PubMed databases by using the keywords: Two investigators independently extracted data and reached a consensus on all of the items. A third investigator took part of data extraction in case of disagreement in any of the items. The data extracted regarded country of origin, first author, and year of publication, number of cases and controls, and genotype frequencies.vs RP + PP and RR + RP vs. PP), homozygous (RR vs. PP), heterozygous (RR vs. RP and RP vs. PP) and the allele (R vs. P) were used. Genotype frequencies were summarized and evaluated by χ2 test of heterogeneity in 2×2 contingency tables with 95% CIs  or heterogeneous (P < 0.05), respectively. The OR and their corresponding 95% CI were used to test the association between the 72 codon polymorphism and breast cancer. All analyses were performed with BioEstat® 5.0 software. To estimate a combined effect, OR were calculated for both fixed and random effect analyses, by applying 95% CIs and individual or combined weights for the studies , Allele-Specific PCR (AS-PCR), Amplifluor®, GoldenGate® Genotyping Assay (GGA), PCR-Denaturing Gradient Gel Electrophoresis (PCR-DGGE), PCR-Restriction Fragment Length Polymorphism (PCR-RFLP), sequencing and Taqman PCR. Breast cancer patients and controls subjects were mainly heterozygous in Asia  and Africa , while the RR homozygous was predominant in America  and Europe . The R allele was predominant in breast cancer patients from America and Europe , Africa (63.2%) and Asia (58.3%).In the last 10 years, eligible studies on R72P polymorphism in the vs. PP and RR vs. RP + PP had OR calculated using a random-effect model. No association between breast cancer risk and the dominant model RR + RP vs. PP  was found  showed a markedly increased risk of breast cancer associated with the RP and PP genotypes, considering the P allele as dominant , as described in Table The P allele has been associated with earlier breast cancer onset in vs. PP , as described in Table Ethnic and geographical nonspecific factors, further to allele frequencies variations in different health populations, have been argued as the reason to the controversial data on R72P role in breast cancer (Dokianakis et al. TP53 gene and the breast cancer risk. The overall data showed an increased risk due to the P allele dominant model, but not in Asia where the risk was associated with the R allele and R dominant model. The present meta-analysis is supported by a large sample size composed by 25,629 cases and 26,633 controls from 41 studies.In conclusion, we found significant association between the R72P polymorphism in the


Transporter associated with antigen processing 1 (TAP1) I333V gene polymorphism has been suggested to be associated with type 1 diabetes mellitus (T1DM) susceptibility. However, the results from individual studies are inconsistent. To explore the association of TAP1 I333V gene polymorphisms with T1DM, a meta-analysis involving 2246 cases from 13 individual studies was conducted. The pooled odd ratios (ORs) and their corresponding 95% confidence intervals (95% CIs) were evaluated by a fixed-effect model. A significant relationship was observed between TAP1 I333V gene polymorphism and T1DM in allelic , dominant , homozygous , heterozygous  and additive  genetic models. No significant association between TAP1 I333V gene polymorphism and T1DM was detected in a recessive genetic model  in the entire population, especially among Caucasians. No significant association between them was found in an Asian or African population. TAP1 I333V gene polymorphism was significantly associated with increased T1DM risk. V allele carriers might be predisposed to T1DM susceptibility. Diabetes mellitus (DM) is a common dysmetabolic syndrome that causes internal insulin insufficiency or insulin resistance, and contributes to the glucose and lipid metabolism disturbance led by the combined action of genetic and environmental factors. Type 1 DM (T1DM), also known as insulin-dependent DM (IDDM), is an organ-specific autoimmune disease characterized by selective destruction of the insulin-producing islet beta cells of the pancreas [Transporter associated with antigen processing (TAP) gene is located in the class II region between DQB1 and DPB1. The TAP1 gene product is a transporter associated with the endogenous antigen processing and presentation. The TAP1 can transport peptides into the endoplasmic reticulum, where they combine with MHC I molecules. The MHC I molecules and antigen peptide compounds are subsequently expressed in the cytomembrane and recognized by the CD8+ T cell receptor, thereby stimulating a cytotoxic lymphocyte protective immune response [et al. found that MHC I molecules are significantly reduced and depleted of CD8+ T cells in transgenic mice with a disrupted TAP1 gene [TAP1 gene deficiency and mutation might lead to the endogenous antigen processing and transportation barrier, which results in such autoimmune disease as T1DM. The 1207th base adenine (A) of the TAP1 gene is substituted by guanine (G), which results in the wild-type isoleucine  being replaced by valine  at 333rd amino acid.Human leucocyte antigen (HLA) region, also known as major histocompatibility complex (MHC) region, located in 6p21, are the major gene regions for T1DM. Approximately 60% of T1DM susceptibility is attributed to HLA gene, in which class II region is most strongly associated with T1DM. A total of 42% of T1DM susceptibility comes from this region. AP1 gene . TAP1 geTAP1 I333V gene polymorphism and T1DM exist, the results from individual studies remain controversial. In 1994, Cucca et al. reported that no primary association exists between TAP1 I333V gene polymorphism and IDDM in Italy [et al. found that no significant difference existed in the distribution of TAP1 I333V alleles between IDDM patients and normal controls, and they concluded that TAP1 I333V gene polymorphism did not exhibit a primary association with Japanese IDDM [et al. determined the relative risk of TAP1 I333V gene polymorphism and T1DM using single-stranded conformation polymorphism in the United States [et al. reported that TAP1 333V allele was the susceptibility allele for T1DM in a Chinese population [Although some studies on the relationship between ese IDDM . In contd States . In 2004pulation .TAP1 I333V gene polymorphism and T1DM.In this study, a meta-analysis involving 1140 T1DM patients and 1108 controls was conducted to determine the relationship between Transporter associated with antigen processing 1’, ‘I333V’, ‘T1DM’ and ‘polymorphism’ were adopted to search the electronic databases of PubMed, Embase, Web of Science, China National Knowledge Infrastructure and China Biological Medicine Database. The last research was updated on 31 October 2013, and the included publication years ranged from 1992 to 2007.The terms ‘i) evaluation of the TAP1 I333V gene polymorphism and T1DM; (ii) T1DM diagnosis and classification criteria were presented by the American Diabetes Association and revised by the World Health Organization in 1997; (iii) case–control or cohort study published in an official journal; and (iv) the study should follow the Hardy–Weinberg equilibrium (HWE).The chosen studies had to be in accordance with the following criteria: , recessive (VV versus IV+II), dominant (IV+VV versus II), homozygous (VV versus II), heterozygous (IV versus II) and additive (V versus I) genetic models, were adopted. The relationship between TAP1 I333V gene polymorphism and T1DM was compared using odds ratio (OR) and its corresponding 95% confidence interval (CI). Chi-squared-based Q-test was used to calculate the heterogeneity between the individual studies, and significance was set at P < 0.05 [P < 0.05.Six genetic models, namely, the allelic , dominant , homozygous , heterozygous  and additive  genetic models. No significant association between TAP1 I333V gene polymorphism and T1DM was detected in a recessive genetic model  in the entire population, especially among Caucasians. No significant association was found between them among the Asian or African population , dominant (Pheterogeneity = 0.007), heterozygous (Pheterogeneity = 0.004) and additive (Pheterogeneity = 0.043) genetic models. The following meta-regression was performed to explore the source of heterogeneity. In the allelic genetic model, the heterogeneity could only be explained by the publication year (P = 0.026). The entire population was separated into two subgroups based on publication year. The manuscripts published before 1995 belonged to subgroup 1, and the manuscripts published after 1996 were classified into subgroup 2. In subgroup analysis stratified by publication year, significantly increased T1DM risk was detected in subgroup 2 , whereas no significant increase in T1DM risk was found in subgroup 1 . Heterogeneity was observed in subgroup 1 , but not in subgroup 2 . These results indicate that heterogeneity mainly existed in the studies published before 1995, and the conclusions were relatively consistent in the studies published after 1996.Significant heterogeneity was observed in the allelic (PP = 0.039), II genotype number of T1DM group sample size , IV genotype number of T1DM group sample size  and IV genotype number of control group sample size . All the individual studies were classified into two subgroups according to IV1. The studies with IV1 < 30 were grouped into subgroup 1, whereas the studies with IV1 > 30 were grouped into subgroup 2. In the subsequent subgroup analysis stratified by IV1, significantly increased T1DM risk was only found in subgroup 2 . In subgroup 1, no significant increase in T1DM risk was detected . Heterogeneity decreased in two subgroups but remained significant .In the dominant genetic model, heterogeneity could be explained by publication year (P = 1.0 × 10−10), IV0 (P = 0.005), total sample size of control group  and weight (P = 0.002). All the individual studies were divided into two subgroups according to II1. Subgroup 1 was defined as II1 < 40, and subgroup 2 was defined as II1 ≥ 40. In the subsequent subgroup analysis stratified by II1, significantly increased T1DM risk was detected in subgroup 1 . In subgroup 2, no significant increase in T1DM risk was found .In the heterozygous genetic model, heterogeneity could be explained by II1 (P = 0.003), VV genotype number of T1DM group sample size  and VV genotype number of control group sample size . Individual studies were separated into two subgroups according to II1, which was similar to that in the heterozygous genetic model. In the following subgroup analysis stratified by II1, significantly increased T1DM risk was detected in both subgroups . The heterogeneity disappeared in both subgroups .In the additive genetic model, heterogeneity could be explained by II1 (TAP1 I333V gene polymorphism and T1DM (OR = 1).Logistic regression was performed on multivariable-adjusted risks, such as age, sex, region and ethnicity. These risk factors had no effect on the association between TAP1 I333V gene polymorphism and T1DM was found in allelic , dominant , homozygous , heterozygous  and additive genetic models . No significant association between TAP1 I333V gene polymorphism and T1DM was detected in a recessive genetic model .In the subgroup analysis stratified by ethnicities among the Caucasian population, a significant association between TAP1 I333V gene polymorphism and T1DM was found in the allelic , recessive , dominant , homozygous  and heterozygous genetic models . A significant association was found between them in a additive genetic model  gene polymorphisms were chosen, constructed and analysed. The three haplotypes, namely, TAP1A (333I-637A), TAP1B (333V-637G) and TAP1C (333V-637A) haplotypes, were constructed and analysed. In the following meta-analysis, the TAP1A haplotype could decrease T1DM risk . The other two haplotypes, namely, TAP1B and TAP1C, had no association with T1DM risk .The funnel plot and Egger*s test were used to evaluate the publication bias of the individual studies. No publication bias was observed by visual inspection of the Begg*s funnel plot in the additive genetic model Fig. . No signTAP1 I333V gene polymorphism and T1DM in the allelic (OR: 1.35), dominant (OR: 1.462), homozygous (OR: 1.725), heterozygous (OR: 1.430) and additive (OR: 1.348) genetic models, especially among Caucasians. No positive association between TAP1 I333V gene polymorphism and T1DM was observed in a recessive genetic model (OR: 1.384), which could be associated with the less VV genotype number in the general population. Table TAP1 333V allele possibly increased T1DM risk, i.e. V allele possibly conferred T1DM susceptibility to humans. No significant association was found between TAP1 I333V gene polymorphism and T1DM in an Asian or African population. The different results from the different populations were probably associated with the different ethnicities. Moreover, the relatively few papers on Asian and African populations possibly led to the negative results, which implied that large-scale studies should be performed in future.In this meta-analysis, a significant relationship was observed between P < 0.05).Given that heterogeneity existed in the allelic and heterozygote genetic models, meta-regression was performed to determine the source of heterogeneity. In the subsequent heterogeneity source analysis, publication year, IV1 and II1 were suggested to be the central confounding factors to explain the source of heterogeneity in allelic, dominant, heterozygous and additive genetic models  was replaced by guanine (G), which resulted in the corresponding amino acid aspartate  being substituted by glycine  at 637th amino acid, of which 637G allele possibly increased the T1DM risk. Thus, analysis of the TAP1 gene 333IV-637AG haplotypes was performed [TAP1A (333I-637A) haplotype had a protective effect for T1DM, which was in agreement with the aforementioned meta-analysis results. The negative association between the TAP1B (333V-637G) and TAP1C (333V-637A) haplotypes and T1DM was possibly associated with the limited number of studies because only five and three papers were included respectively. Thus, the results need to be further verified by more studies.Transporter associated with antigen processing 1 gene polymorphisms result in different recognition and transport affinity to the same endogenous antigen peptide, and cause different bodies to show various immune responses to the same endogenous antigen, including protective immunity, immune tolerance and autoimmune tendency [TAP1 I333V gene polymorphism and T1DM remains unclear. In 1998, Quadri et al. reported that TAP1 333I allele specifically enhances translocation of model peptides containing basic C-terminal amino acid residue. However, TAP1 333V allele did not show specificity for the peptide with basic amino acid residue C-terminus, which indicated that the TAP1 I333V mutation could lead to the change in specificity of the transported peptides. Some peptides transported by mutated TAP1 gene products might cause the expression deficiency of MHC I molecules, which resulted in the spontaneous termination of immune tolerance, launched the autoimmunity process and caused the destruction of pancreas islet β cells [TAP1 333V allele increased the T1DM risk, which could be associated with the differences in peptide transport. However, this result is still a hypothesis that needs confirmation by future experiments.tendency . The mec β cells . In the et al. found that the IV genotype of TAP1 gene locus 333 was a susceptible gene for T1DM that had a possible function in the production of autoantibodies, such as insulin autoantibody, islet cell antibody and glutamic acid decarboxylase antibody. Thus, the autoimmune response and destruction in the pancreas islet β cells resulted in T1DM. However, the concrete mechanisms by which the TAP1 333 IV genotype promoted autoantibody production remained unclear [In 2004, Sheng  unclear .TAP1 I333V gene polymorphism are still insufficient. The TAP1 expression level was influenced not only by TAP1 I333V gene polymorphism but also by other genetic and environmental factors, such as ethnicity, inflammation state and other immune system diseases. Given that T1DM is a multigenic heredity disease, TAP1 I333V gene polymorphism might be associated with the gene linkage disequilibrium, such as TAP1 Val458Leu, Asp637Gly and Arg648Gln, which increase T1DM susceptibility [The following limitations were noted in the present meta-analysis. Large-scale studies on the association of T1DM with tibility .Transporter associated with antigen processing 1 I333V gene polymorphism was positively associated with increased T1DM risk. Patients with the V allele may be susceptible to T1DM. Our results may help in establishing individual T1DM therapy strategies. In consideration of the limitations, more large-scale studies are needed to elucidate the significance of our conclusions.


Genetic polymorphism was hypothesized to be reason of variation in prostate cancer incidence among different racial group. Based on that published data on the association of prostate cancer susceptibility with polymorphisms in genes encoding Glutathione S-transferases (GSTs) were inconclusive, the aim of this study was to more precisely address the role of GSTs polymorphisms  on prostate cancer risk in Asian descent.A meta-analysis including 8 articles with 711 cases and 1122 controls for GSTT1 and 1098 cases and 1588 controls for GSTM1 was performed.Significantly increased prostate cancer risk was found among subjects carrying GSTM1 null genotype (odds ratio (OR) = 1.403; 95% confidence interval (CI) = 1.088 – 1.808) but not among subjects carrying GSTT1 deletion genotype . When stratified by country, the null genotype of GSTT1 neither increased nor decreased prostate cancer risk significantly in China , Japan , and Korea . While significant association of elevated prostate cancer risk with GSTM1 deletion were found in China  and Korea  but not in Japan .In summary, this meta-analysis suggested that the null genotype of GSTM1 rather than GSTT1 may be involved in the etiology of prostate cancer in Asian population. Prostate cancer is the most commonly diagnosed malignancy and the second leading cause of cancer-related deaths in the western male population . In AsiaLow-penetrance susceptibility genes combining with environmental factors may be crucial in the development of cancer . PolymorThe roles of GSTT1 and GSTM1 polymorphism in prostate cancer risk have been studied previously, but most published results are limited to western populations. Up to now, several original studies have reported the relationship between these polymorphism and prostate cancer risk in men of Asian descent –16; howeAs showed in Figure 2 = 58.9%; p value = 0.045). It's clear that this analysis is based on pooling data from different country. When subgroup analyses were conducted in different country's group, the null genotype of GSTT1 also neither increased nor decreased the risk of prostate cancer significantly in China, Japan, and Korea .After all the extracted data were pooled, a total of 711 cases and 1122 controls were available for GSTT1 analysis. As detailed in Table 2 = 60.7%; p value = 0.013). Then, analyses stratified by different country showed a significant association of elevated prostate cancer risk with GSTM1 deletion in China and Korea, the ORs were 1.665 (95%CI = 1.324 – 2.094) and 1.914 (95%CI = 1.311 – 2.793) respectively. However, such significant connection between GSTM1 null genotype and increased prostate cancer risk were not present in Japan subjects .The data of 1098 cases and 1588 controls derived from pooled data were usable for meta-analysis of GSTM1. In Table 2 = 0.0%, p value for heterogeneity = 0.626). For GSTM1 null genotype, when deleting the study of Komiya et al [2 = 30.1%; p value = 0.198). Importantly, statistically significant association of increased prostate cancer risk with GSTM1 deletion was still not substantially altered . These evidences indicated that the present results were statistically robust.For GSTT1 deletion genotype related to prostate cancer susceptibility, several studies , 13 mighya et al  based onPublication bias of literatures was evaluated using Begg's funnel plot and Egger's test. The shapes of the funnel plots for GSTT1 and GSTM1 seemed to be symmetrical, but the extent of symmetry couldn't be obtained quantificationally. Therefore, Egger's test was applied to provide statistical evidence of funnel plot symmetry, and the results still did not reveal any evidence of publication bias Figure –3b.The present meta-analysis from 8 studies mainly evaluated the potential role of GSTT1 and GSTM1 polymorphisms in prostate cancer susceptibility in Asian population.Variation in prostate cancer incidence among different racial groups and geographic locations has been well documented , 4. For The current study through meta-analysis indicated that significantly elevated prostate cancer risk was associated with GSTM1 rather than GSTT1 null genotype. As statistically significant between-study heterogeneity was found when carrying out meta-analysis for both GSTT1 and GSTM1, subgroups analyses in the different country were conducted. Then we found similar non-significant relationship between GSTT1 deletion and prostate cancer risk among in china, Korea, and Japan, while significant association of null genotype of GSTM1 with increased prostate cancer risk was detected in China and Korea but not in Japan. When studies led to heterogeneity were removed according to sensitivity analysis, the pooled ORs of both GSTT1 and GSTM1 were not substantially altered, indicating that the present results were statistically robust. These results were similar to some previous studies and simultaneously inconsistent to other studies , 24. TheIn the GST family, GSTM1 has been proved to directly implicate in DNA adduct formation caused by benzo(a)pyrene, the main component of cigarette smoke ; GSTT1 iThe limitations of this meta-analysis should be acknowledged. Firstly, only published studies were included while studies with null or unexpected results had a lower probability of being published and weren't included in meta-analysis, thus bias may have appeared. Secondly, all of studies included in our systematic analysis were the hospital-based studies which inherently have selection biases because such controls may not represent the general population, particularly when the genotypes under investigation were associated with the disease-related conditions that the hospital-based controls may posses. Thirdly, heterogeneity among studies was existed, which may be derive from the study design, the source of controls, the differences of genetic background, and the environment presented among different country. Fourthly, the shortage of funnel plots for evaluating publication bias was not avoided, as the reported reasons that the probability of exactly detecting publication bias through funnel plot was equivalent to a coin toss  and the Despite some limitations, this meta-analysis suggested that the GSTM1 null genotype was associated with enhanced risk of prostate cancer carcinogenesis, but non-significant association was observed for GSTT1 deletion with prostate cancer risk. However, high-quality epidemiological studies using large sample, standardized unbiased genotyping methods, homogeneous prostate cancer patients and well matched controls and considering gene-gene and gene-environment interactions would be necessary to conduct in the future. Such studies taking these factors mentioned above into account might eventually provide our better, comprehensive understanding of the association of the polymorphism of GST family (including GSTT1 and GSTM1) with prostate cancer susceptibility in Asian even worldwide population.A comprehensive systematic search through Medline, PubMed, Embase, and Web of Science was conducted for all medical publications up to May, 2014. The search strategy was performed using the search terms “prostate”, “polymorphism”, “Glutathione S-transferase”, “GST”, “Glutathione S-transferase T1”, “GSTT1”, “Glutathione S-transferase M1” and “GSTM1” without language restriction. After all searched studies were retrieved, the reference lists of their papers and review articles were checked and hand-searched to ascertain additional undetected but eligible studies. Only published studies with full-text papers, rather than any meeting or conference abstract, were included. When more than one of the same patient population was included in several articles, only the most recent or complete study was used in this meta-analysis.The following inclusion criteria were considered: (a) case-control or cohort studies using Asian population that had original data for a quantitative evaluation of the association of GSTT1 and/or GSTM1 present/null polymorphism with prostate cancer; (b) an appropriate description of GSTT1, and GSTM1 polymorphisms in cases and controls; (c) cases with prostate cancer were eligible regardless of whether they had a first-degree relative with prostate cancer or not, regardless of tumor stage; (d) controls were eligible if they were male, with or without BPH, or other diseases; (e) sufficient published data for calculating an odds ratio (OR) with 95% confidence interval (CI).The exclusion criteria were: (a) studies without the raw data of genotype of GSTM1, and GSTT1; (b) case-only studies, family-based studies, case reports, editorials, and review articles ; (c) studies that compare the racial variation of GSTT1 and/or GSTM1 variants in healthy population; (d) controls with any type of tumor; (e) studies that used GSTT1 and/or GSTM1 polymorphisms to predict survival in prostate cancer; (f) studies that investigated GSTT1 and/or GSTM1 variants as makers for response to therapy.According to the inclusion/exclusion criteria mentioned above, information was carefully extracted from all eligible articles independently by two of the authors. Any disagreement was resolved by discussion between the two authors. If these two authors could not reach a consensus, another author was consulted to resolve the dispute and a final decision was made by the majority of the votes. For each study, the following data were reviewed and collected: first author’ name, year of publication, the country of origin, ethnicity of the study population, subject source, involved genes, total number of cases and controls, and numbers of cases and controls with the present/null genotypes. Subject source were classified to population-based studies and hospital-based studies. We did not define any minimum number of patients to include in this meta-analysis.p value > 0.10 for the Q-test indicated a lack of heterogeneity among studies, so the pooled OR estimate of each study was calculated by the fixed-effects model (the Mantel-Haenszel method)[p value less than 0.05 from t-test suggested by Egger was considered representative of statistically significant publication bias [The strength of relationship between the GSTT1 and/or GSTM1 present/null polymorphism and prostate cancer susceptibility was evaluated using crude ORs with 95%CI. The pooled ORs were performed for null versus present genotype. Heterogeneity assumption was tested by the chi-square-based Q-test . A p vall method). Otherwil method). The stal method). An estiion bias . All the


Background. The associations between polymorphisms in microRNAs and the susceptibility of colorectal cancer (CRC) were inconsistent in previous studies. This study aims to quantify the strength of the correlation between the four common polymorphisms among microRNAs  and CRC risk. Methods. We searched PubMed, Web of Knowledge, and CNKI to find relevant studies. The combined odds ratio (OR) with 95% confidence interval (95% CI) was used to estimate the strength of the association in a fixed or random effect model. Results. 15 studies involving 5,486 CRC patients and 7,184 controls were included. Meta-analyses showed that rs3746444 had association with CRC risk in Caucasians . In the subgroup analysis, we found significant associations between rs2910164 and CRC in hospital based studies . rs2292832 may be a high risk factor of CRC in population based studied . Conclusion. This meta-analysis showed that rs2910164 and rs2292832 may increase the risk of CRC. However, rs11614913 polymorphism may reduce the risk of CRC. rs3746444 may have a decreased risk to CRC in Caucasians. Colorectal cancer (CRC) is the third leading cause of cancer death for both men and women in the USA . In EuroHowever, the mechanism of colorectal carcinogenesis is still not fully understood. Compared with other complex diseases, CRC may be caused by both genetic and environmental factors . BecauseMicroRNAs (miRNAs) are a large class of small noncoding RNAs, which participate in various biological processes and may regulate tumor suppressor genes or oncogenes . Single Relevant articles were independently searched by two authors (Wang M and Dai ZJ) in PubMed, Web of Science, and Chinese National Knowledge Infrastructure (CNKI). The keywords were as follows: colorectal cancer/colorectal carcinoma, microRNA 146a/149/196a2/499, and polymorphism/genotype/variant/SNP. All qualified studies prior to December 20, 2014, were included. The eligible literature must be published in English or Chinese. Furthermore, reference lists of main reports and review articles were also reviewed manually to identify additional relevant publications.The following criteria were used to select eligible studies for further meta-analysis: (1) case-control design; (2) full-text study; (3) studies that evaluated the associations between miRNA polymorphisms and CRC risk; and (4) studies that included detailed genotyping data.Articles were reviewed independently by two reviewers and data with discrepancies were discussed by all authors. For each included study, the following information was collected: first author, year of publication, country of origin, ethnicity, source of control, total number of cases and controls, and genotyping methods as well as number of cases and controls with the different genotypes. Different ethnic groups were categorized as Caucasian, Asian, African, and “mixed.” All the case and control groups were well controlled.Z test.The associations between miRNA polymorphisms and CRC risk were measured by odds ratio (OR) with 95% confidence interval (CI). The significance of the pooled OR was determined by the Q and I2 statistics. If the P value of heterogeneity test was more than 0.05 (P > 0.05), the pooled OR estimate of the study was calculated by the fixed-effects model. Otherwise, the random-effects model was used. The value of the I index was used to assess the degree of heterogeneity . Publication bias was evaluated by the funnel plot. All statistical analyses were carried out with the review manager version 5.1 .The meta-analysis assessed association by using 5 different genetic models: homozygous genetic model, heterozygote genetic model, dominant genetic model, recessive genetic model, and allelic model. Hardy-Weinberg equilibrium (HWE) among the control subjects was tested by the Chi-square test. Statistical heterogeneity among studies was assessed with the According to the inclusion criteria defined above , fifteenAs shown in http://dx.doi.org/10.1155/2015/276410. There were 11 studies with 3,937 cases and 5,120 controls for rs2910164. miR-146a rs2910164 polymorphism has no association with CRC risk in the overall population . After omitting the study which was not according with the HWE, the results were in accordance with the overall population . When stratifying analysis by ethnicity, there was also no significant association observed between the rs2910164 and CRC susceptibility in the five genetic models. In the analysis stratified by the source of control, significant associations were observed in the hospital based studies for recessive genetic model . No associations were found in population based studies.The main results of this meta-analysis were presented in Table S1, in Supplementary Material available online at P = 0.36, P = 0.44; CT versus CC: OR = 1.35, 95% CI = 0.64–2.86, P = 0.43; TT versus CC + CT: OR = 0.91, 95% CI = 0.54–1.53, P = 0.71; TT + CT versus CC: OR = 1.24, 95% CI = 0.82–1.87, P = 0.31). And there was also no association existing in any genetic models after we rejected the studies which were not in agreement with HWE. There was only one study based on Caucasian. When excluding the Caucasian study, the null association remained in Asians. We further made stratified analysis based on the source of control. And we observed significant associations in population based studies in the recessive genetic model .The association of the rs2292832 polymorphism with CRC susceptibility was investigated in 5 studies with 1,568 cases and 1,824 controls. We failed to find any significant associations in any genotype . Further subgroup analysis by ethnicity showed no association between rs11614913 and CRC risk either in Caucasians or in Asians. And no significant associations were observed in population based studies and hospital based studies for all genetic models.Ten studies with 2,906 cases and 4,150 controls were used to evaluate the relationship between rs11614913 polymorphism and CRC risk. No significant association was detected under all the genetic models . Further analysis of the studies which were in agreement with HWE also showed no association between the rs3746444 polymorphism and CRC. And null associations were found in hospital based studies or population based studies. However, in the stratified analysis according to ethnicity, the homozygote model demonstrated a significant decrease in the CRC risk in Caucasians  and CRC risk. In this study, we found that rs3746444 polymorphism has no association with CRC risk in the overall population. However, accumulative data from two studies based on Caucasian background (317 cases and 477 controls) showed significantly decreased cancer risks in the homozygous model. Considering the limited sample size, we need more large well-designed studies to evaluate the result. Simultaneously, we failed to find any significant correlation between the rs2910164, rs2292832, and rs11614913 polymorphisms and risk of CRC in overall studies or in different ethnic groups for all genetic models. But significant association between rs2910164 and CRC was observed in hospital based studies in the recessive model. And we found that rs2292832 polymorphisms may be associated with increased the risk of CRC in population based studies for recessive model. While T allele of rs11614913 was a protect factor of CRC in the homozygous model when we excluded the studies which were not in agreement with HWE.Although the biological mechanism of SNPs in miRNAs contributing to the regulation of cancer susceptibility and development remains unknown, a large quantity of meta-analyses has been engaged in the newly developed field in the past few years . A previSome limitations still existed in this meta-analysis. Firstly, the included studies are mainly based on Asian background. There were only five studies based on Caucasian background and no studies on African background. Secondly, some detailed information  was not considered. Further large scale multicenter studies based on Caucasian or African will be needed to clarify the possible roles of these polymorphisms in CRC. Though many investigators have been devoted to identify the roles of the miRNAs in carcinogenesis, the mechanisms were still unclear. Further biologically functional studies are warranted to explain the molecular mechanisms.In summary, this meta-analysis showed that there were no associations between the four polymorphisms and CRC risk in the overall population. Further analysis stratified by the source of control showed that rs2910164 polymorphism is associated with CRC risk in hospital based studies, and rs2292832 polymorphism may contribute to the susceptibility of CRC in population based studies. In the pooled analysis from all studies which are in agreement with HWE revealed that rs11614913 TT carriers may have a decreased CRC risk. Moreover, in the stratified analysis according to ethnicity, rs3746444 polymorphism may reduce the risk of CRC in Caucasians.Table S1: Meta-analysis results.


The null association persisted when the analysis was stratified by sex or restricted to non-smokers. In the stratification by study location, the summary OR for the studies from China and for those from Japan was 0.77 (95% CI = 0.60–0.99) and 1.21 (95% CI = 0.94–1.54), respectively (P for differences = 0.04). Cumulative epidemiologic evidence suggests that green tea consumption is not associated with pancreatic cancer.Emerging laboratory and animal studies indicate that green tea inhibits development and progression of pancreatic cancer, but evidence from epidemiologic studies appears inconsistent and inconclusive. A meta-analysis summarizing published case-control and cohort studies was performed to evaluate the association of green tea consumption with risk of pancreatic cancer. Pertinent studies were identified by a search of PubMed and EMBASE up to April 2014. A random-effects model was assigned to compute summary risk estimates. A total of three case-control studies and five prospective studies were included, comprising 2317 incident cases and 288209 subjects. Of them, three studies were from China and the reminders were conducted in Japan. Overall, neither high  Pancreatic cancer, which is currently the fourth leading cause of cancer death in men and women, is an aggressive malignancy characterized by poor prognosis and patient survival, with a 5-year survival rate of less than 5% . DespiteGreen tea is one of the most popular beverages in Asia, particularly in China and Japan. Over the past 2 decades, potential chemopreventive effects of green tea constituents on cancer development have drawn substantial attention because of a large body of evidence from laboratory and animal studies continuously demonstrating a favorable effect of green tea on cancers of different sites, including pancreatic cancer ,4,5,6. NAll case-control and cohort studies published in English-language journals from 1966 to April 2014, reporting the association between green consumption and risk of pancreatic cancer were identified by searching PubMed and EMBASE using the search terms “green tea”, in combination with “pancreatic” or “pancreas”. The citation sections of recovered articles were also reviewed to identify additional studies. To be included, studies also had to report relative risks (RRs) or odds ratio (OR) with 95% confidence intervals (CIs) of pancreatic cancer associated with green tea consumption. For each study, the following characteristics were collected: author, publication year, country of origin, study design, sex of subjects, sample size, method for exposure measurement, the RRs or ORs and 95% CIs that reflected the greatest degree of control for potential confounders, and variables accounted for in the analysis. Study selection and data were carried out independently by two authors.OR with 95% CI is the common measure of the association in this meta-analysis, and RR in the cohort studies were considered approximately as ORs. The results that were reported by sex separately were combined with a fixed-effects model, and the combined results were used in the meta-analysis. The random-effects model accounting for both within- and between-study variation was assigned to compute the summary risk estimates . Subgrouet al. [A dose-response analysis according to the method proposed by Greenland and Longnecker  and Orsiet al.  was alsoet al. ,11 that I2 statistics [P-value of less than 0.1 was considered statistically significant heterogeneity; and for the I2 statistic, we set the following cut-off points: <25% (low heterogeneity), 25%–50% (moderate heterogeneity) and >75% (severe heterogeneity). Potential publication bias was evaluated using Egger’s test and Begg’s funnel plot [Statistical heterogeneity was assessed using Q and atistics . For thenel plot . All staThe search strategy yielded 288 citations , of whicet al. [et al. [et al. [The reported findings on the association of green tea consumption and risk of pancreatic cancer among individual studies varied substantially. An overall significant/non-significant inverse association were reported in three studies ,11,14 . Results of Egger’s test indicated no evidence of publication bias (P = 0.31), which could be supported by a visual examination of Begg’s funnel plot (not shown). P for interaction = 0.04).Adjustment for cigarette smoking is important because smoking often highly correlated with tea consumption, especially among Asian population. Restricting the analysis to the studies that took into account smoking, the summary OR was 0.93 (95% CI = 0.74–1.18). Three studies ,12,16 alP = 0.04, I2 = 56.9%).There were six studies ,13,14,15To our knowledge, the present study is the first quantitative analysis of the association between green tea consumption and risk for pancreatic, based on published results from 8 observational studies involving more than 2300 cases. Overall, our findings suggest that green tea consumption is not associated with risk of pancreatic cancer. Interestingly, a 33% reduction in risk associated with high green tea consumption was observed in Chinese population. Tea polyphenols, especially (−)-epigallocatechin-3-gallate (EGCG), which is an active ingredient in green tea, have been shown to have antimutagenic, antigenotoxic, and anticarcinogenic properties . It is sThe differences between experimental and epidemiologic findings may be attributable to several reasons. On one hand, the amount of green tea consumed in human studies may not be large enough to reach an effective dose inhibiting pancreatic carcinogenesis because the doses of tea polyphenols used in some of experimental studies are much higher than those observed in human plasma and tissue after green tea consumption . On the When stratifying the analysis by geographic area, we found a significant inverse association of green tea consumption with risk of pancreatic cancer among Chinese population, but a positive (non-significant) one among Japanese. Similar results for colorectal cancer have been reported in a recent meta-analysis . DiffereP for differences = 0.50). A pooled analysis of three studies [Smoking has been suggested to be an important factor confounding the association of tea consumption and cancer risk because green tea drinkers, in particular male drinkers were more likely to be smokers ,11. Poss studies ,16 of noThe relatively low incidence of pancreatic cancer compared with other common cancers results in small number of incident cases in cohort studies. Therefore, individual studies have been underpowered to examine this relation across a broad range of green tea consumption. This meta-analysis quantitatively summarizing cumulative evidence from published cohort and case-control studies comprising more than 2300 cases largely enhanced the statistical power. However, except for possible influences of residual confounding and exposure misclassification mentioned above, several other limitations should also be acknowledged. First, the levels and the ranges of green tea consumption, as well as statistical adjustment in primary studies differed substantially, which may have contributed to heterogeneity amongst studies. Second, it is possible that cancer patients might have altered their recent tea drinking habits as a result of their pre-clinical illness, which may bias our findings. Three cohort studies ,13,14 haIn summary, findings from the present meta-analysis indicate that green tea consumption is not associated with risk of pancreatic cancer. The results of lower risk associated with high consumption among Chinese population were mainly derived from case-control studies, and so warrant further confirmation.