TAP2) has been identified to play an important role in the HLA-associated diseases and immune response. The goal of our meta-analysis was to summarize the contribution of TAP2 polymorphisms to the risk of RA.Rheumatoid arthritis (RA) is a severe chronic immune mediated inflammatory disease that has been shown to be associated with human leukocyte antigen (HLA) loci. The transporter associated with antigen processing 2 , TAP2-565Ala > Thr (rs2228396) and TAP2-665Thr > Ala (rs241447). The meta-analyses were involved with 9 studies  among 973 cases and 965 controls.Meta-analyses were performed between RA and 3 TAP2-379Ile allele was significantly associated with an increased risk of RA  = 1.44, 95% confidence interval (CI) = 1.18-1.74). This association was further shown only in the dominant model . Subgroup analyses by ethnicity revealed that the association of TAP2-379Ile was significant in Asians . In addition, another significant association of TAP2-565Thr allele with RA was observed in Europeans .Meta-analyses showed that TAP2-379Ile allele was significantly associated with a 59% increased risk in the dominant effect model. Subgroup analyses by ethnicity showed that TAP2-379-Ile increased the risk of RA by 38% in Asians and TAP2-565Thr increased the risk of RA by 38% in Europeans.Our meta-analyses suggested that http://www.diagnosticpathology.diagnomx.eu/vs/2097080313124700The virtual slide(s) for this article can be found here:  Rheumatoid arthritis (RA) is a common immune-mediated chronic inflammatory disease . Severe TAP2) gene encodes transporter 2, ATP-binding cassette, sub-family B (MDR/TAP) that is a major histocompatibility complex (MHC) gene located between HLA-DP and HLA-DQ [TAP2 has been shown to play an important role in the HLA-associated diseases and immune response [Consistent association was found between human leukocyte antigen (HLA) loci and RA . The trad HLA-DQ . TAP2 ded HLA-DQ , especiad HLA-DQ . TAP2 haresponse .TAP2 polymorphisms have been tested for their association with the occurrence and development of RA [TAP2 polymorphisms with RA, yielding only 2 significant results  without time and language restriction, using the keywords “rheumatoid arthritis TAP2 association” and “rheumatoid arthritis TAP2 polymorphism”. The obtained studies would be included in our meta-analyses if they met the following criteria: (1) It was an original case–control study with an assessment of the association of TAP2 with RA in humans; (2) It contains sufficient information to infer the odds ratios (ORs) and 95% confidence intervals (95% CIs); (3) Genotype distribution of each polymorphism in controls met Hardy-Weinberg equilibrium (HWE); (4) The cumulative number of individual studies for one genetic locus are at least three. We extracted or calculated the following information from each selected study: the first author, year of publication, country, ethnicity, genotyping method, numbers of cases and controls, control source, HWE for controls, reported association results, power of each involved study and minor allele frequency (MAF) in each stage.We collected studies that examined the associations of TAP2-379, TAP2-565, and TAP2-665) [2 test [2 < 50%), and the random-effect model would be used to the studies with significant heterogeneity (I2 > =50%). Combined ORs and CIs were estimated by Review Manager 5 [Since some studies presented the data of haplotypes  ,26. We uAP2-665)  to test  [2 test  were useanager 5 . Funnel TAP2 gene were initially collected. Of them, we excluded 7 studies that were not related to RA, 1 case-only study, 4 studies without genotyping information, 1 study that did not meet HWE. Finally, 9 articles [TAP2 polymorphisms . Subgroup analysis by ethnicity showed that this significant association was only found in Asians . A further subgroup meta-analyses by ethnicity showed that significant heterogeneity was coming from the Europeans for TAP2-379 (I2 = 64%) and Asians for TAP2-565 (I2 = 76%). Further subgroup studies showed that significant heterogeneity also existed in the meta-analyses of TAP2-565  and TAP2-665 .Significant statistical heterogeneity was found in the meta-analyses of All the power analyses in current meta-analyses were tested under a moderate risk of SCZ  with dominant effect such as -607A/C polymorphism of -18 gene , rs1048923R gene , -173G/CMIF gene  and -607-18 gene . Our obsTAP2 polymorphisms in Asians and Europeans separately. For TAP2-379Ile, significant association was found in Asians but not in Europeans. On the contrary, we observed TAP2-565Thr as a risky factor of RA in Europeans but not in Asians. This might be due to a lack of power in the subgroup meta-analysis in Europeans for TAP2-379  and in Asians for TAP2-565 . To be noted, one Asian study [TAP2-665Ala was found a high allele frequency (allele frequency = 0.601, Table Since MAFs are different in different populations, we further evaluated the role of an study  involvedTAP2 polymorphisms were only evaluated in Europeans and Asians. The findings might not be feasible for other populations such as Africans. And potential difference in intra-European and intra-Asian population might influence the result of our study (see the MAFs in Table TAP2 polymorphisms and RA. Fourthly, certain multiple testing existed in the current study, and cautions needed to be taken for the significant results. Fifthly, there are 2232 polymorphisms in TAP2 gene according to the NCBI dbSNP database. Our study only focused on three polymorphisms of TAP2 that might be hard to give fully consideration of the contribution of TAP2 polymorphisms. Moreover, the 3 TAP2 polymorphisms might not be the causal variants but be in high linkage disequilibrium with other established RA MHC variants. Sixthly, since significant associations were found in the both allelic and dominant model for TAP2-379, our results implied the genetic models of TAP2-379 polymorphism were complex and hard to be determined. Seventhly, there were publication biases for the current meta-analyses of TAP2-379 and TAP2-565. After removing the outlier studies [Our meta-analyses presented several limitations that needed to be taken with cautions. Firstly, the associations of  studies ,17, the TAP2-379Ile allele was significantly associated with a 59% increased risk in the dominant model. To be more specifically, TAP2-379-Ile increased the risk of RA by 38% in Asians and TAP2-565Thr increased the risk of RA by 38% in Europeans. Future large-scale and well designed studies are required to confirm our findings and to reveal other TAP2 polymorphisms with contribution to RA disease.In summary, our meta-analyses suggested that RA: Rheumatoid arthritis; SNPs: Single nucleotide polymorphisms; HWE: Hardy-Weinberg equilibriu; ORs: Odds ratios; 95% CIs: 95% confidence intervals.None of the authors have any commercial or other association that might pose a conflict of interest. All authors are responsible for the content and writing of the paper.SD participated in research design. YC, PR, XZ, YH and LX selected the articles, QH, LT, GP, DL, QG and YL performed data analysis. The manuscript was drafted by DD, YC and SD, and critically reviewed and discussed with the other co-authors. All the authors read and approved the final manuscript.Dongjun Dai and Yong Chen: Co-first authors of this work.


P=8.8 × 10−5; using data from 34,840/114,981 T2D cases/controls and 63,746/130,681 CHD cases/controls). FG in non-diabetic individuals tends to increase CHD risk (OR=1.15 (1.00–1.32), per mmol·per l, P=0.05; 133,010 non-diabetic individuals and 63,746/130,681 CHD cases/controls). These findings provide evidence supporting a causal relationship between T2D and CHD and suggest that long-term trials may be required to discern the effects of T2D therapies on CHD risk.In observational studies, type-2 diabetes (T2D) is associated with an increased risk of coronary heart disease (CHD), yet interventional trials have shown no clear effect of glucose-lowering on CHD. Confounding may have therefore influenced these observational estimates. Here we use Mendelian randomization to obtain unconfounded estimates of the influence of T2D and fasting glucose (FG) on CHD risk. Using multiple genetic variants associated with T2D and FG, we find that risk of T2D increases CHD risk (odds ratio (OR)=1.11 (1.05–1.17), per unit increase in odds of T2D,  In order to effectively design interventions, it is useful to understand the complex interplay between multiple syndromes. Here, Ahmad et al. use genome-wide association study data and Mendelian randomisation to examine the influence of Type 2 diabetes and fasting glucose levels on coronary heart disease. Understanding the role of type-2 diabetes (T2D) in the pathogenesis of coronary heart disease (CHD) is a fundamental problem for the design of effective approaches for preventing cardiovascular disease22Despite this observational evidence, recent large-scale randomized controlled trials (RCTs) have shown conflicting and inconclusive results on the effect of intensive glucose-lowering therapies on the short-term prevention of CHD in patients with T2D. Four large-scale RCTs56711This contradictory evidence has led to the suggestion that elevated blood glucose does not cause CHDConfounding may strongly influence observational studies, particularly when confounding factors are unknown or inadequately measured11Mendelian randomization (MR) is a study design in which genetic variants are employed as instrumental variables for estimating the unconfounded effect of an exposure  on a disease 2122In the present study, we analyse summary-level genome-wide association study (GWAS) data from multiple genetic variants to obtain MR estimates of the effect of T2D and FG on CHD. We assess FG in non-diabetic subjects since FG in the non-diabetic range has previously been associated with CHD risk2425P<5 × 10−8 for the allelic effect of each SNP on T2D risk) associations with T2D in the largest GWA study to date HNF1B [TCF2] locus), neither the index SNP nor any of its proxies (defined as variants with linkage disequilibrium r2>0.9 in HapMap CEU population) was found in either the CARDIoGRAMplusC4D or CARDIoGRAM dataset. This variant was thus excluded from further analyses. The set of 37 T2D risk-increasing variants for which both T2D and CHD data were available constituted the set of candidate variants, and SNPs in this set were further evaluated in our MR study  using cross-phenotype meta-analysis (CPMA), a statistical procedure for using summary-level GWAS data to identify pleiotropic associations among traitsP=1.7 × 10−4 for MR analyses; I2=63% (95% CI: 47–74%)  and two with large negative effects  on CHD risk. Because of their pleiotropic associations, they were excluded from MR analysis ;  47–74%) . Of the analysis . As discP=8.8 × 10−5, I2=38 (95% CI: 1–62%)  , Fig. 2.: 1–62%) .BCAR1 with a large effect on T2D risk dominated the heterogeneity in our MR estimate; removing this variant produced a risk estimate of markedly lower heterogeneity: (OR=1.09 (95% CI: 1.05–1.14); P=5.0 × 10−5 for MR analysis; I2=15% (95% CI: 0–48%)). Two other variants in the MR-instrument set (at the KLDHC5 and CDKN2A/B loci) were found through exhaustive subgroup analysis to make important contributions to the heterogeneity in the MR estimate. Removing from the MR-instrument set the three variants associated with the BCAR1, KLHDC5 and CDKN2A/B loci yielded a risk estimate with a heterogeneity estimate of zero. (OR=1.12 (95% CI: 1.07–1.16); P=2.7 × 10−7 for MR analysis; I2=0% (95% CI: 0–45%)).To assess whether individual variants made large contributions to the heterogeneity of the MR estimate, we exhaustively computed and compared random-effects estimates of all subgroups of non-pleiotropic variants in the MR-instrument set, using a previously described combinatorial approachGCKR, IRS1, KLF14 and PPARG) also influenced additional metabolic traits such as LDL cholesterol (N=8) yielded an MR estimate of the effect of T2D on CHD risk of 1.07 (95% CI: 1.01–1.14); P=0.02 for MR analysis; I2=0% (95% CI: 0–68%)) and estimates based on T2D risk alleles with no clear effect on the above mechanisms (N=18) also conferred increased risk of CHD (OR=1.13 (1.04–1.23); P=3.4 × 10−3 for MR analysis; I2=60% (32–76%)) ) .P<5 × 10−8) associations with FG levels (−1 increase in FG (95% CI: 0.021–0.035 mmol l−1) (We identified 33 variants that showed genome-wide significant (G levels  using daG levels . Data onmol l−1) .Just as for T2D risk alleles, we used the CPMA statistic to test the FG candidate variants for pleiotropic associations with LDL-C, triglycerides, systolic blood pressure, diastolic blood pressure and BMI . Of thes−1 increase in FG (95% CI: 1.04–1.54 CHD odds per 1 mmol l−1 increase in FG); P=0.02 for MR analysis; I2=39% (7–60%) (−1 FG (95% CI: 1.00–1.32 CHD odds per 1 mmol l−1 increase in FG); P=0.05 for MR analysis, I2=0% (0–44.6%)  . The eff0–44.6%) .To assess whether related glycemic traits in non-diabetics also were causally associated with increased risk of CHD, we undertook a similar analysis of HbA1c levels on CHD risk, which did not yield statistically significant results .Using summary-level data for T2D and FG risk alleles, our MR study supports observational evidence, suggesting that T2D and FG lead to CHD. These findings have important implications both for the care of T2D patients, and for the design of clinical trials to assess the effect of T2D treatments on cardiovascular risk; they support the hypothesis that lowering T2D risk and glucose levels can help prevent CHD. While these findings contrast with recent short-term RCTs investigating the effect of glucose lowering in T2D567567Importantly, almost all current treatments for T2D focus on lowering glucose levels. Our study considered the effects of FG on CHD for individuals without T2D. However, since the effect of FG on CHD risk is considerably higher among diabetic individualsA relatively small proportion (4.8%) of the variance in FG levels in the non-diabetic population is explained by the common genetic variants used in this studyA strength of our study is that data on associations between exposure, outcome and confounder traits were typically gathered in different population samples; this approach reduces the possibility of over-fitting effect-size estimates. Moreover, the fact that we draw effect-size data from separate large-scale GWA studies for exposure and outcome traits means that effect sizes are more precisely assessed than would be possible by the analysis of individual-level data from a smaller study.Although this has the potential to introduce error because variation in risk exposure between study populations can distort the estimated effect, this error is expected to be small since the various studies that were part of the DIAGRAMv3, CARDIoGRAM and MAGIC meta-analyses often drew subjects from the same overall population samples. Several factors can lead to bias in our estimate of the effect of genetically elevated risk of T2D and FG223334MR has previously been used to show that non-FG in non-diabetic individuals is causally related to ischaemic heart diseaseIn addition to non-fasting blood glucose, numerous other traits have been assessed through MR studies for their impact on cardiovascular risk3647In this study, we used MR analysis of summary-level GWAS data to provide evidence that genetically increased risk of T2D leads to increased CHD risk. We also provide evidence through MR analysis for a trend, indicating that increases in FG in non-diabetics leads to an increase in CHD risk. Our results support the hypothesis that the relationship between T2D and CHD is indeed causal, and that associated metabolic traits in non-diabetics may also have a causal influence on CHD risk. These findings suggest that long-term efforts to prevent T2D and lower glucose levels can decrease CHD risk.P value≤5 × 10−8) SNPs associated with increased T2D risk identified in the DIAGRAM consortium, the largest meta-analysis to date of the T2D GWAS studiesWe gathered data from large meta-analyses of GWA studies examining the exposure (T2D and FG), outcome (CHD) and confounder traits  from the largest GWAS studies to date2122We gathered GWAS data on FG from the largest meta-analysis to date (carried out by the MAGIC consortium)For each of the susceptibility variants for T2D and FG, we sought summary-level data for CHD from the CARDIoGRAMplusC4D Metabochip study, since this is the largest GWAS meta-analysis for CHD to dateIn all, 37 of 38 significant T2D variants were represented either in the CARDIoGRAMplusC4D or CARDIoGRAM datasets. One of the 38 T2D variants, namely rs11651052, was absent from both the CARDIoGRAMplusC4D and the CARDIoGRAM datasets. Moreover, no variant in close linkage disequilibrium with rs11651052 could serve as a proxy for it in our analysis. For this reason, we excluded it from further analysis. The remaining set of 37 candidate variants provided the basis for our analysis of the effect of T2D on CHD, and contains most of the lead variants used previously in genetic risk scores for predicting the risk of T2D refs . All 33 For pleiotropic traits, summary-level results were similarly sourced from the largest GWAS conducted to date for each trait: (i) LDL-C5455P values called CPMAP values across phenotypes to that predicted by the null hypothesis of no pleiotropic association, under which P values are uniformly distributed. Variants with any detectable association with pleiotropic traits were removed from the analysis, and the remaining non-pleiotropic variants were taken as instruments for the MR analysis  of ui on the exposure x and outcome y, respectively, and let s(βi) denote the s.e. of βi. Then the MR estimate associated with the ith genetic variant is:For each instrument, we obtained an estimate of the effect of the exposure on the outcome using summary-level data. Let and the variance of this estimate is:ith MR estimate of α by wi=1/vi. The inverse-variance-weighted fixed-effects estimate is then:Define the precision of the s(αfixed) of this estimate is given byand the s.e. αfixed may also be interpreted as the regression coefficient resulting from the generalized linear regression of the outcome effect size βi on the exposure effect size γi assuming heteroskedastic errors; in this regression, the ith error term has a variance equal to s(βi)2, and the offset coefficient in the regression is zero.We observe that αrandom estimate and its s.e. s(αrandom) are constructed from the individual estimates using standard methodsI2, which reports the fraction of the total variance in the meta-analytic estimate that is due to intrinsic variability in the effect size, as distinct from the variability arising due to measurement errorαrandom and its s.e. s(αrandom) are given by equations analogous to those for αfixed and s(αfixed), in which the weights assigned to individual estimates are adjusted to take into account heterogeneity in the effect size.The random-effects estimate For each of the exposure traits (T2D and FG), we carried out a meta-analysis of estimates obtained from individual susceptibility variants using both fixed-effects and random-effects models to obtain pooled estimates of the effect of exposure on outcome. Such meta-analytic methods are commonly used to summarize information from independent studies for the effect of an intervention on a health outcome. In this study design, we use independent genetic variants as instruments to assess the effect of exposure (T2D or FG) on outcome (CHD), and then pool these individual estimates using statistically efficient estimators formally analogous to those of inverse-variance-weighted meta-analysisFor each of the exposure traits, we also carry out a random-effects meta-analysis. Because the heterogeneity in the effect sizes for individual SNPs is large for each of the exposures, our random-effects estimates are close to an unweighted average of the effect sizes.(αfixed) for the fixed-effects model and by exp(αrandom) for the random-effects model.For all MR meta-analyses, we report estimates from the random-effects models  had pleiotropic effects .Supplementary Figures 1-5, Supplementary Tables 1-7, Supplementary Notes 1-3 and Supplementary References


Autism spectrum conditions (ASC) are a group of neurodevelopmental conditions characterized by difficulties in social interaction and communication alongside repetitive and stereotyped behaviours. ASC are heritable, and common genetic variants contribute substantial phenotypic variability. More than 600 genes have been implicated in ASC to date. However, a comprehensive investigation of candidate gene association studies in ASC is lacking.In this study, we systematically reviewed the literature for association studies for 552 genes associated with ASC. We identified 58 common genetic variants in 27 genes that have been investigated in three or more independent cohorts and conducted a meta-analysis for 55 of these variants. We investigated publication bias and sensitivity and performed stratified analyses for a subset of these variants.P values below a threshold of significance of 0.01. Of these 15 variants, 11 were re-investigated for effect sizes and significance in the larger Psychiatric Genomics Consortium dataset, and none of them were significant. Effect direction for 8 of the 11 variants were concordant between both the datasets, although the correlation between the effect sizes from the two datasets was poor and non-significant.We identified 15 variants nominally significant for the mean effect size, 8 of which had This is the first study to comprehensively examine common variants in candidate genes for ASC through meta-analysis. While for majority of the variants, the total sample size was above 500 cases and 500 controls, the total sample size was not large enough to accurately identify common variants that contribute to the aetiology of ASC.The online version of this article (doi:10.1186/s13229-015-0041-0) contains supplementary material, which is available to authorized users. Autism spectrum conditions (ASC) are a group of neurodevelopmental conditions characterized by difficulties in social interaction and communication alongside unusually repetitive and stereotyped behaviour and unusually narrow interests . ASC hasSequencing and copy number variation analyses have identified a number of rare, highly penetrant, possibly causative variants. Strategies to identify common variants through genome-wide association studies have failed to produce consistent, replicable results across cohorts . This maIn the field of psychiatric genetics, studies have comprehensively investigated existing candidate gene studies and used meta-analysis to investigate genetic associations –10. In thttps://gene.sfari.org/) and HuGE Navigator (http://hugenavigator.net/). Since both of these databases do not completely document the available literature, we additionally searched PubMed, Scopus and Google Scholar. The search terms used were ‘Gene name’ or ‘variant ID’ and ‘Autism’ or ‘Autistic Disorder’ or ‘Asperger Syndrome’.A preliminary literature search of genes associated with ASC was performed using SFARI gene  they reported effect sizes or statistics to measure effect sizes and confidence intervals; (2) the studies were either a case-control association study or a transmission disequilibrium study of autism; (3) the variants did not deviate from Hardy-Weinberg Equilibrium (HWE) in the control group or if the sample size was too small to effectively calculate HWE due to sampling effect. Though we checked for HWE in family-based studies, this was not a requirement for including these studies as the study design overcomes the issue of population stratification; (4) cases had a diagnosis of an autism spectrum condition  according to DSM-IV, DSM-5 or ICD-10 criteria; (5) the global minor allele frequency (MAF) of the variant investigated was greater than 0.01; (6) the studies were reported in English and (7) the common variants were investigated in independent cohorts. Authors of the articles were contacted if sufficient information was absent to use the data for meta-analysis. In addition to the published studies, we used unpublished genotype data from two cohorts from our research group at the Autism Research Centre, University of Cambridge. These cohorts are labelled ‘Chakrabarti ’ and ‘Wa they repP values <0.05 we calculated fail-safe N.Meta-analysis was performed only if variants were investigated in three or more independent cohorts. Family-based association tests (FBATs) studies were not included as effect sizes are not calculated in FBA. For variants investigated in five or more independent cohorts, we performed a complete meta-analysis. This included the calculation of effect size and publication bias, sensitivity analysis and subgroup analysis. For variants investigated in three to five independent cohorts, we performed a partial meta-analysis restricted to the calculation of mean effect size. We did not perform a meta-analysis for variants investigated in fewer than three cohorts as there was insufficient power to significantly investigate the underlying effect. For variants with I2 statistics in conjunction with Q-statistics. A fixed effect model was applied if the P value for Q-statistics was above 0.05 and I2 was below 60. The random effects model was used if either the P value was below 0.05 or I2 was above 60, as an I2 above 60 indicates that 60 % of the total observed variation is due to true heterogeneity [All analyses were performed using Comprehensive Meta-Analysis version 2.0 . Meta-anogeneity , 10.P values <0.05, we computed both classic fail-safe N and Orwin’s fail-safe N to check the number of studies required to make the P value non-significant and make the effect size trivial respectively. For Orwin’s fail-safe N, the non-significant odds ratio (OR) was kept at 1.05 or 0.95 depending on the effect direction. While this is certainly not a trivial effect size, it is difficult to identify variants with such small effects with precision given the sample sizes in the meta-analysis. Subgroup analysis was performed after stratifying based on ethnicity or study methodology to check if either of these variables affected the final effect size. We conducted the subgroup analysis only for variants investigated in five or more independent cohorts. Meta-analysis was performed only if there were at least three independent cohorts after stratification to account for power considerations.Egger’s regression in conjunction with a funnel plot was used to assess publication bias. Sensitivity analyses were performed by removing each study from the meta-analysis and calculating the mean effect size for the remaining studies. This analysis was used to assess the contribution of each study to the final weighted effect in the analysis. Additionally, for the variants with OR and 95 % confidence intervals (CI) were used to calculate the mean effect size. For transmission disequilibrium tests (TDT), odds ratios were calculated according to methods laid out by Kazeem and Farall . Where pAge was not regarded a confounding variable as ASC is a neurodevelopmental condition, and genetic variations are largely invariant across lifespan. However, ASC has a male-female ratio of 5:1 . We did P values for variants with P values <0.05 in our meta-analysis. The autism PGC dataset is the largest available and accessible GWAS dataset for autism. The sample size of any of the variants investigated through meta-analysis in the study, except rs4141463 in MACROD2, is smaller than the sample size of the PGC autism dataset. Despite this, the PGC dataset is underpowered to detect variants with small effects. We were motivated to investigate the top variants in our study in the PGC dataset to ascertain if the candidate variants were at least nominally significant (P < 0.05) and if the effect direction was concordant between the two samples.While we did not choose to include data from available GWAS due to potential overlap of participants, we compared the results using the publicly available GWAS dataset from the Psychiatric Genomics Consortium (PGC). In the ASC cohort of the PGC dataset, 4788 trio cases and 4788 trio pseudocontrols as well as 161 cases and 526 controls have been genotyped. Details of the cohort, genotyping methods and statistical analysis are given elsewhere . We searWe identified 463 genes that have been tested for genetic association using HuGE Navigator (as of August 2014). SFARI Gene reports 616 genes to be associated with autism (as of August 2014). Only 185 of these genes have been examined in ASC using genetic association studies. Of these, we identified 89 genes from the SFARI Gene list that were not included in the HuGE Navigator list, bringing the total list of potential genes to 552. We did not identify any additional genes from AutismKB database. Thus, we reviewed 552 genes in total for the meta-analysis.Scopus, Google Scholar and PubMed were searched for publications relating to ASC and any of the 552 genes. We searched for common variations in these genes that have been investigated for ASC in at least three independent cohorts. Using the eligibility criteria outlined in the methods section, we identified 27 genes that could be taken forward for meta-analysis. In total, there were 58 common variants across these 27 genes that were investigated in our meta-analysis. Details of the studies included and excluded for the 27 genes are given in Additional file OXTR [RELN [SLC6A4 [HOXA1 [HOXB1 [MTHFR [. Detailed information of previous meta-analyses is provided in Additional file HOXB1, STin2 VNTR in SLC6A4 and the GGC repeat in RELN. These three variants were excluded from the current meta-analyses as we could not identify additional data to add to the original meta-analyses. For the sake of comprehensiveness, we have included the data for these three variants in our table. Of the remaining 55 variants, we conducted a complete meta-analysis for 20 variants and a partial meta-analysis for 35 variants. A flow chart of the study protocol is given in Fig. We next searched the literature for existing meta-analyses for the 58 variants and 27 genes in ASC, identifying existing meta-analyses for OXTR , RELN [1TR [RELN , SLC6A4  [SLC6A4 , HOXA1 [4 [HOXA1 , HOXB1 [1 [HOXB1  and MTHF1 [MTHFR . DetaileP values below 0.05. Nine of these variants had a P value below 0.01. The most significant association was rs167771 in DRD3 . Seven other significant associations with P values <0.01 were in CNTNAP2 , RELN , OXTR , SLC25A12 , EN2  and MTHFR . As expected for common variants in ASC, the odds ratios for the alleles tested were small and lay between 0.781 (0.446–1.368) for MAOA uVNTR and 1.822 (1.398–2.375) for DRD3 rs167771. Details of the variants analysed, model used and the P values are provided in Table Effect sizes for 15 variants in 12 genes had P values below 0.05. Of these, the most significant three variants  were also significant in the non-stratified analyses. Stratification did not increase the significance for these variants. A variant in EN2 (rs1861973) was significant after stratifying based on both ethnicity (Caucasian only) and study methodology (TDT). Another variant in EN2 (rs1861972) was significant after stratifying for study methodology (TDT). Finally, the STin2 variant in SLC6A4 also exhibited a significant trend in the Caucasian-only subgroup. This result indicates that at least for a few variants implicated in ASC, ethnicity and study methodology can potentially influence the outcome. Results of the subgroup analyses are provided in Table We performed subgroup analyses, stratifying by ethnicity and study methodology, for variants originally investigated in five or more independent cohorts. In the stratified analyses, six variants had OXTR (Egger’s test (two-tailed) P value = 0.03). However, the mean effect size for the variant was not significant . Notably, sensitivity was significant for some variants. Of the nine variants with P values below 0.01, we performed sensitivity analyses on the six variants with data from more than five independent cohorts . For rs1801133, most studies contributed approximately equally, with the exception of two studies [P value of the OR . The latter variant was analysed using a fixed effect model and becomes non-significant when analysed using a random effect model. For the remaining two variants (rs1861972 in EN2 and rs362691 in RELN), the classic fail-safe N is above 10. The sample sizes, however, are modest. These analyses indicate that the first two variants are likely to be false positives. With rs1861972, the significance in P value is driven largely by the TDT-only subset in the original analysis , and the two variants are separated by 152 base pairs. In this study, we used the random effects model to meta-analyse rs1861973 and it was not significant. Stratifying by both study methodology and ethnicity reduced the heterogeneity considerably, allowing us to use a fixed effect model. For rs1861973, both a Caucasian-only and a TDT-only subset were significant  had data only from three studies and had a relatively high OR of 1.82 to 1.40–2.38. The small sample size for this variant inflated the OR making it significant. The effect direction was discordant for the variant in the PGC dataset, and it was not significant in this dataset.While the sample sizes for most variants were competitive for candidate gene association studies , these are not sufficient to accurately calculate effect sizes. Additionally, the different study methodologies and ethnicities contributed to heterogeneity in the sample which potentially confounded the analyses. It is clear from this study that significant heterogeneity exists for a large fraction of the variants tested. In fact, heterogeneity is significantly and positively correlated with the number of independent datasets included per variant in the analyses, indicating that the current study may not have uncovered all the heterogeneity. We were able to remove some of the heterogeneity after stratifying for ethnicity and study methodology, but heterogeneity influenced the results for some for the variants even after this. This indicates that other additional factors contribute to variance in the effect. One potential source of heterogeneity is finer population stratification. Fine-scale population stratification cannot be addressed in candidate gene association studies as these test only a few variants. Further, HWE which is used to check for population admixture among other issues is performed individually for each variant in these studies thereby failing to utilize multi-marker information to correct for population stratification. We were unable to stratify based on sex or clinical ascertainment two factors known to contribute to heterogeneity in ASC. It is unclear how clinical heterogeneity maps onto genetic heterogeneity in ASC. Existing genetic studies that stratify based on IQ or other clinical phenotype and subphenotypes have had limited success , 25. TheRELN [SHANK3 [GABRB3, GRIN2B and SCN2A. However, there was not enough evidence to evaluate the role of common variants in ASC for these genes through the current meta-analysis.Another cause for concern is the small number of genes with enough data to meta-analyse. Of 552 genes, we had data for only 27 of these, less than 5 %. None of the 27 genes analysed were ASC risk genes as predicted by DAWN . Further [SHANK3 , none of [SHANK3 . A few gThe majority of the studies analysed were of Caucasian ethnicity. We were able to stratify for a Caucasian ethnicity for some of the variants, but were not able to stratify for other ethnicities due to power considerations. It is also noteworthy that the PGC autism dataset used a Caucasian sample for analyses, and to our knowledge, there is no well-powered GWAS that investigates the role of common variants in autism in other ethnicities. Since the minor allele frequencies of the alleles tested and the variants tagged by these allele can vary depending on ethnicity, this makes it difficult to compare the results of the non-stratified meta-analyses with the PGC autism dataset. Replicating the top variants in well-powered samples from different ethnicities will help understand the ethnicity-specific risk for each variant.The candidate gene association studies typically have small samples, which overestimate effect sizes. The lack of replication do not indicate that these loci do not contribute to the aetiology of ASC, but, rather, that there is insufficient evidence to implicate it in ASC. ASC is highly polygenic, and more than 49 % of its heritability can be attributed to common variants . As effeen masse, contribute to a significant fraction of ASC, there have not been any sufficiently powered studies to date to identify important common variants. We attempted to address this issue using a meta-analysis of candidate gene association studies. Though this is the first comprehensive study of candidate gene association studies in ASC, it failed to identify causative variants—11 of 15 variants with P values <0.05 were not significant in a larger sample from the PGC. Data was unavailable for the remaining five variants in the PGC dataset. We discuss the potential issues with such an approach and underline the need for much larger sample sizes to accurately identify common variants that contribute to ASC.While recent studies , 3 have


ESRα, PvuII (rs2234693) and XbaI (rs9340799) were studied while in ESRβ gene, risk of infertility was determined for silent G/A RsaI (rs1256049) polymorphism. Total 124 subjects (74 cases and 50 controls) were part of this study having primary infertility. Restriction fragment length polymorphism (RFLP) was performed with PvuII, XbaI and RsaI to determine polymorphism. Correlation between age and follicle stimulating hormone (FSH) of cases and controls was determined and no association was found between infertility and FSH hormone. Heterozygous AG genotype of XbaI polymorphism (P= 2.505e-06) and heterozygous TC genotype (P= 0.00003) in PvuII polymorphism were strongly associated with risk of infertility. In ESRβ gene, there was lack of polymorphism for RsaI in our population as all subjects were homozygous (GG). Haplotype frequencies showed that XbaI and PvuII polymorphisms are in strong linkage disequilibrium. This study shows that in our population XbaI and PvuII polymorphisms of ESRα are associated with risk of infertility.Studies on polymorphism of estrogen receptor (ESR) alpha and beta genes have been mostly implicated in infertility, but the results have been controversial due to lack of comprehensive data. The present study focused on association of ESR genes with both male and female infertility. In  Infertility has become a common problem as it is affecting almost 1 in 6 couples. It is defined as the incapability to achieve fertilization after a certain period of time following sexual intercourse (mostly 1 year or 6 months in certain cases) without any contraceptive measures . InfertNumerous genes are involved in the complex process of reproduction. A qualitative diagnosis of infertility could only be done by focusing on female and male physical abnormalities, endocrine irregularities and genetic conditions of both partners associated with reproduction .ESRα, ESR1) and β  genes are involved in controlling the physiological response to estrogen. ESRα gene is present on chromosome 6q25. It consists of 8 exons which are parted by 7 intronic regions having a complete size of 140 kb  and luteinizing hormone (LH) which consequently leads to deficient or absence of puberty . Estrogal., 1988). The ESal., 1997), while PvuII (rs2234693) and XbaI (rs9340799). These are found in the intron 1 and divided by only 46 bp. The PvuII (T397C) polymorphism occurs due to T/C transition in first intron, while the XbaI (G351A) polymorphism is caused by a G/A transition located 50 base pairs downstream of the PvuII polymorphic site  and AluI (rs4986938) . Severa RsaI rs16049 and ESR1 XbaI polymorphism was suggested to have an effect on azoospermia or idiopathic severe oligospermia while in women ESR1 PvuII strongly affected pregnancy rate after in vitro fertilization (IVF) . PvuII al., 2007) and conRsaI polymorphism (rs1256049) with male infertility has been established in Caucasian patients. It is located in exon 5 of ERβ  on the ESRα and ESRβ polymorphisms in the etiology of male and female infertility. To the best of our knowledge, there is no study so far in international literature, associating ESRα and ESRβ polymorphisms with both male and female infertility.The aim of this study is to determine the importance of For polymorphic analysis of ESRα and β gene, total of 124 individuals  were included in this study. Blood samples of 74 infertile patients were collected from LIFE . Every participant has completed detailed questionnaire, regarding their medical and surgical history, lifestyle habits, exposure to gonadotrophins and family history. The control group consisted of 50 subjects . Males had normal semen parameters i.e. sperm count over 15 million sperm/ml according to WHO guideline  of whicEach subject donated 5 ml of blood for genomic DNA extraction and serum analysis. 2 ml of each peripheral blood sample was transferred to a sterile EDTA coated vacuotainer  and remaining 3 ml was transferred into sterile tubes without having anti-coagulant for separation of serum . Serum was separated immediately after sample collection and stored at -20 °C until hormone profiling whereas blood vials were properly labeled and stored at -20 °C. An inclusion and exclusion criteria was set to carefully evaluate the patients in order to identify the exact cause and to select only primary infertile people. All male patients underwent semen analysis according to WHO's recommendations  in respSerum FSH was measured by a microplate immune enzymatic assay using immunoassay kit, catalog number: BC-1029 . Standard protocol as mentioned in the kit was used for all samples.® Genomic DNA Purification Kit, (Promega) was used. Manufacturer recommendations were followed for isolation of DNA.For isolation of genomic DNA from whole blood WizardPvuII (c.454 -397 T > C) and XbaI (c.454 -351 A>G) polymorphisms, PCR amplification of the polymorphic regions was performed using oligonucleotide primers as described elsewhere  and c.454-351 A/G (XbaI) polymorphisms, digestion of the amplified PCR fragment of 1372 bp was performed with PvuII (Fermentas#ER0631) and XbaI (Fermentas#ER0681) restriction enzymes, separately. The reaction mixture was mixed gently and spins down for few seconds. Then the reaction vials were kept at 37 °C water bath for 1 hour and stored at -20 °C until visualization on ethidium bromide stained 2.0 % agarose gel by gel doc.For determination of ERα genotype, polymerase chain reaction restriction fragment length polymorphism (PCR-RFLP) analysis was performed. For determination of al., 2010). For PCRsaI (1082 G>A) polymorphism in ERβ gene was identified by PCR amplification and restriction enzyme digestion. Total volume of reaction mixture for PCR amplification was 25 µl consisting of 12.5 µl of master mix (fermentas #K0171), 2 µl (32 pmol) of each forward and reverse primer, and 2 µl of template DNA. For restriction digestion, the amplified fragment was digested with RsaI (Fermentas#ER1121). The reaction mixture was mixed gently and spins down for few seconds. Then the reaction vials were kept at 37°C water bath for 1 hour and stored at -20°C until visualization on ethidium bromide stained 2.0 % agarose gel. The primers, PCR conditions, restriction enzymes, and DNA fragment sizes are listed in Table 1The ERα and ERβ genes polymorphisms and the risk of infertility was also evaluated using Odds Ratio at 95 % confidence intervel. Haplotype construction and haplotype frequencies were determined by using the Haploview software version 3.2 available at http://www.hapmap.org. Linkage disequilibrium analysis was also performed with the Haploview program. The level of significance was set at P < 0.05. Data were analyzed with SPSS 16.0 for Windows  software.The significance of differences in the genotype distribution and allele frequency were tested using the χ2 test, which was also used to determine the degree of subjects conforming to Hardy-Weinberg equilibrium, single genotype and the frequency of allele between each group. Clinical variables were compared using the independent sample t-test. An Odds Ratio (OR) at 95 % confidence intervals (CI) was calculated with respect to the presence of the reference genotype. The association between the combined genotypes of the Classifications of patients on basis of types of infertility are described in Table 2P=0.016)  are given in Table 3PvuII polymorphism, there was significant deviance from HWE for cases  but no observable deviance from HWE for controls  was determined. For XbaI polymorphism, there was evident deviance from HWE for both cases  and controls   Table 3.XbaI and PvuII polymorphism with the risk of infertility. Odds Ratio (OR), 95 % Confidence interval (C.I.) and χ2 values were determined  and G (p =0.83783) of cases and controls with risk of infertility. However heterozygous AG genotype showed strong association with risk of infertility in cases when compared to homozygous AA (p=2.505e-06) as well as rare homozygous GG genotype of controls. Rare homozygous GG (p=0.97111) and homozygous AA (p=0.97111) genotype of cases showed no significant association with risk for infertility. Individuals with genotype of AG and of GG (p=0.00069) in cases showed significant association with risk of infertility. This showed strong association of heterozygosity with risk of infertility in our population  were ped Table 4. The sign Table 4. PvuII polymorphism, although there was difference between T and C allele frequencies in cases and controls (Table 4p=0.76013). The heterozygous TC genotype of PvuII showed more prevalence in infertile group and was significantly associated with risk of infertility when compared with homozygous TT (p=0.00003) and rare homozygous CC (p=0.00005) genotype of cases and controls. Homozygous TT (p=0.84154) and rare homozygous CC genotype of cases showed no significant association with risk of infertility. Allele positivity was determined by determining the number of individuals possessing a particular genotype in cases and control group. Infertile subjects with heterozygous genotype TC when combined with infertile subjects having rare homozygous genotype CC (p=0.00158), showed significant association with risk of infertility  . All the cases and controls were analyzed for this polymorphism, but no polymorphism was detected in any subject (both patients and controls). All the subjects were homozygous for common allele GG. This indicates absence of this polymorphism in our studied population.In the present study 74 patients (cases) with primary infertility were selected while 50 healthy fertile individuals were considered as controls. In infertile male cases, azospermia was the most frequent cause of infertility (43 %) while in infertile females the major cause of infertility was tubal factor (24 %) and sec tubal factor (23 %). These results correspond to former epidemiological studies which have shown that in a sub fertile population male factor infertility has been observed in 25 %-35 % couples. About 14-22 % have tubal factor, 10-27 % have ovulatory dysfunction, 5-6 % have endometriosis while 10-17 % have unexplained causes of infertility . Correlation between age and FSH of patients and controls was calculated. No correlation was found between age of male and female patients and their endocrine levels. In our work lack of correlation between age and serum FSH levels of cases reveals that the study subjects did not had gonadotrophins related infertility. However a positive collinear relationship was found between the age and FSH of female controls. The increase in basal FSH can be considered as the earliest endocrine marker of reproductive aging . XbaI and PvuII polymorphism of the ESRα gene and RsaI polymorphism of ESRβ gene with fertility status was analyzed. Previous studies on the influence of the ERα and ERβ genes involved in male factor infertility have yielded conflicting results  but there was no observable deviance from Hardy-Weinberg equilibrium for controls . Salanti et al. . Assessal. (2005) had sugal. (2005), for caXbaI polymorphism, there was observable deviance from Hardy-Weinberg equilibrium for both cases  and controls . Allelic frequency for all groups was also calculated and it conformed to Hardy-Weinberg equilibrium. Previous investigations have revealed that in approximately 10 % of gene-disease association studies, there is significant deviation from HWE in genotype frequencies of healthy controls. Only 5 % of these are expected by chance alone even at significance level of P=0.05 with sufficiently powered studies as most of these studies use small underpowered sample sizes .XbaI polymorphism, insignificant association was found between allele A (p=0.83783) and G (p=0.83783) of cases and controls with risk of infertility. However heterozygous genotype AG (p=2.505e-06) showed strong association with risk of infertility in cases when compared to homozygous AA as well as rare homozygous GG genotype of controls. Rare homozygous GG and homozygous AA (p=0.97111) genotype of cases showed no significant association with risk for infertility.In case of XbaI polymorphism and chances of infertility. However in their population the GG genotype  showed about sevenfold increased predisposition to infertility compared to individuals carrying homozygote AA, whereas insignificant disposition was found in heterozygous AG  individuals. They also detected a significant difference in allele frequencies of XbaI polymorphism in cases and controls . However in their population the homozygous AA genotype of XbaI polymorphism was associated with risk of infertility due to its prevalence in infertile group as compared to controls (p=0.014). Allele positivity was determined by determining the number of individuals possessing a particular genotype in affected and control group. Individuals with genotype of AG and of GG  in cases showed significant association with risk of infertility. This showed strong association of heterozygosity with risk of infertility in our population for XbaI polymorphism. These results are in contradiction to the study by Ayvaz et al.  on infePvuII polymorphism, although there was difference between T and C allele frequencies in cases and controls, but this difference was statistically insignificant (p =0.76013). The heterozygous TC genotype of PvuII was more common in infertile group and was significantly associated with risk of infertility when compared with homozygous TT (p=0.00003) and rare homozygous CC (p=0.00005) genotype of cases and controls. Homozygous TT (p =0.84154) and rare homozygous CC genotype of cases showed no significant association with risk of infertility. Individuals for determination of allele positivity with heterozygous genotype TC when combined with cases having rare homozygous genotype CC (p=0.00158), showed significant association with risk of infertility. Thus in our population individuals with heterozygous TC genotype are more predisposed to infertility as compared to individuals with CC or TT genotype. These results varied from those reported in other populations. In Turkish population, for PvuII polymorphism, a detectable association was found between fertile and infertile group for heterozygous TC and rare homozygous CC genotype  . Thus the association observed in these polymorphisms and infertility could be due to only one of them. As a multifactorial condition like infertility is dependent on several polymorphisms, so a third polymorphism in strong linkage disequilibrium with XbaI and PvuII can be the actual cause of this association. The genomic organization of ESRα gene promoter is very complex. Alternative splice sites are present in multiple promoter regions which results in formation of different estrogen protein transcripts. This difference in expression patterns can affect the function and production of ESR protein. Hence the possibility of these polymorphisms affecting the correct splicing of RNA and producing alternatively spliced mRNA variants cannot be ruled out  RsaI polymorphism all the cases and controls were analyzed, but no polymorphism was detected in any subject (both patients and controls). All the subjects were homozygous for common allele GG. This indicates absence of this polymorphism in our studied population.For XbaI and PvuII polymorphisms with risk of infertility. Also detection of significant deviance from Hardy-Weinberg raised several possibilities about this study and the population from which it is derived. Pursuing this course can lead to more useful information regarding ESR gene. More thorough examination of ESR locus is necessary to determine whether other regions of ESRα and ESRβ may also have functional significance related to infertility.In conclusion this work showed significant association between The authors acknowledge the Department of Microbiology and Molecular Genetics, University of the Punjab, Lahore, Pakistan, which provided the facilities to carry out this research. The authors would also like to thank the participants’ families for their support and Hameed Latif hospital for collection of blood samples.None declared.


We conducted a meta-analysis to assess the association between patatin-like phospholipase domain-containing 3 (PNPLA3) rs738409 polymorphism and nonalcoholic fatty liver disease (NAFLD) and its subtypes simple steatosis(SS) and nonalcoholic steatohepatitis (NASH). The study-specific odds ratios (ORs) and 95% confidence intervals (CIs) were calculated using fixed-effects or random-effects models, with assessment for heterogeneity and publication bias. Twenty-three case-control studies involving 6071 NAFLD patients and 10366 controls were identified. The combined results showed a significant association between NAFLD risk and the rs738409 polymorphism in all genetic models . In addition, evidence indicated that the rs738409 polymorphism was significantly associated with NASH in all genetic models . The subgroup and sensitivity analyses showed that these changes were not influenced by the ethnicities and ages of subjects or by the source of controls. The rs738409 polymorphism was only significantly associated with risk of simple steatosis in the allele contrast and had no effect in the other genetic models. These findings suggest that the rs738409 polymorphism in PNPLA3 gene confers high cross-ethnicity risk for NAFLD and NASH development. Nonalcoholic fatty liver disease (NAFLD) is the most common cause of liver disease in western countries, affecting up to 20–35% of the general population2589The human patatin-like phospholipase-3 (PNPLA3) gene is localized on human chromosome 22. The PNPLA3 protein, which is also known as adiponutrin, is expressed in both adipocytes and hepatocytes14151718There is no approved therapy for NAFLD, and the diagnosis of NASH can only be proven by liver biopsy. In addition, it is important to establish whether the associations differ between different subgroups of NAFLD. To clarify the association between the rs738409 polymorphism and risk of NAFLD, we conducted a systematic review and meta-analysis of the available prospective studies with the specific aims of analyzing NAFLD subgroups, including simple steatosis or NASH, to clarify whether the association differed by histological parameters.We conducted an electronic search of the PubMed, EMBASE and Web of Science databases from their inception until December 2014 to identify the association between the rs738409 polymorphism and NAFLD risk using the following search terms: PNPLA3 and (polymorphism or variant or variation) and  or (fatty liver) or steatohepatitis). Additional studies not captured by our database search were identified by surveying the references of the originally identified reviews and research reports and by using the MEDLINE option “Related Articles”. The search was confined to human studies without country restrictions. In addition, the publication language was restricted to English.PNPLA3 rs738409 polymorphism and risk of NAFLD; (2) case-control studies based on unrelated individuals; (3) studies in which the diagnosis of NAFLD was clear; (4) studies that provide the number of NAFLD cases and controls and the frequency of the rs738409 genotypes; and (5) studies published in English. The major reasons for study exclusion were the following: (1) case-only study or overlapping data; (2) studies with a sample size less than one hundred; (3) studies with abstracts only and reports published as comment and review papers; and (4) studies with secondary causes of steatosis, including alcohol abuse, the use of drugs, surgical procedures and hepatitis B and hepatitis C virus infection.Potentially relevant studies were selected based on the following inclusion criteria: (1) studies concerning the association between the Two investigators independently selected the trials and extracted the data, and disagreements or uncertainties were resolved by consensus. The following data were extracted: first author, publication year, country of origin, ethnicity of studied population, sex ratio, mean age, diagnostic criteria for NAFLD, number of individuals in the case and control groups, frequency of PNPLA3 genotypes in the cases and controls; and consistency with the Hardy-Weinberg equilibrium(HWEs).The quality of the studies was assessed independently by two investigators according to the quality assessment scores developed from the genetic association studies conducted by Thakkinstian et al. The total scores ranged from 0 (worst) to 13 (best)2) were used to calculate the heterogeneity among the studies, and a P value < 0.10 or I2 > 50% was considered to be significantThe strength of the association between the PNPLA3 polymorphism and NAFLD risk was assessed by the odds ratios (ORs) and 95% confidence interval (CI). The Chi-square test was used to assess the Hardy-Weinberg equilibrium (HWE) in order to analyze the genotype distribution in the control groups. Meta-analyses were performed for four genotype contrasts per outcome: allele contrast (G versus C), dominant model (GG+CG versus CC), recessive model (GG versus CG+CC), and additive model (GG versus CC)21Three subgroup analyses were additionally carried out by ethnicity , mean age (pediatric or adult) and source of the controls . The statistical analysis was performed with RevMan software version 5 (Cochrane Collaboration) and STATA software version 10.0 (Stata Corporation). A P value < 0.05 was considered to be statistically significant in this trial unless otherwise specified.The search strategy initially identified 419 potentially relevant articles, and 363 articles were determined to be irrelevant after a review of the titles and abstracts. Thus, 56 trials proceeded to a full-text review, and an additional 33 studies were excluded. Finally, 23 articles were ultimately selected for inclusion in the meta-analysis1718192627282930313233343536373839404142434417192627282930313218333435363738394043441618283340414217192629303132343536373839441617181926282930313234353637383942443340411718192627293132333435363738394041424344282 = 89%, P < 0.00001); dominant model ; recessive model ; and additive model   . After e0.00001) .Subgroup analyses were conducted to explore the differences between ethnicity, mean age and sources of the controls. In the subgroup analysis by ethnicity, significant association was found between the rs738409 polymorphism and NAFLD risk among the Caucasian, Asian and Hispanic populations. The association between rs738409 polymorphism and NAFLD was most significant in Hispanic population, which followed by Caucasian population, and the association was weakest in Asian population. The analyses also showed that the risk of NAFLD was significantly increased in both adult participants and pediatric subjects. In addition, the G allele was strongly associated with NAFLD susceptibility in hospital-based controls and population-based controls. The results were consistent in all genetic models. More details are presented in 2 = 0%, P = 0.674) with obvious publication bias (Egger test: P = 0.980)  and the publication bias was not significant (Egger test: P = 0.457)  . The six= 0.457) .2 = 0%, P = 0.58), dominant model and recessive model   . The assSensitivity analysis was performed under additive model to evaluate the influence of a specific study on the overall estimate. The corresponding pooled ORs with 95% CIs produced similarly before and after omitting each study at a time, indicating that our results were stable and reliable . The funThe current meta-analysis provided a systematic assessment of the association between the PNPLA3 rs738409 polymorphism and susceptibility to NAFLD, including its subtypes simple steatosis and NASH. Our results suggested that rs738409 polymorphism exerted a significant influence not only on NAFLD risk, but also on histological severity of NAFLD. In addition, a further analysis showed that individuals with the rs738409 polymorphism experienced a significantly increased risk for NASH. However, our meta-analysis did not show a definite association of rs738409 polymorphism with simple steatosis.Our results are consistent with those from a previous meta-analysis conducted by Sookoian et al.A large population-based study that involved 9229 multiethnic population, including African-Americans, Hispanics and European-Americans, revealed that patients with the rs738409 polymorphism are associated with a higher risk of NAFLD compared with normal controls2738It should be noted that rs738409 polymorphism was only significantly associated with increased simple steatosis risk under allelic model, but not under the other three genetic models. When stratified by ethnicity, we only detect a significant association in the Asian subgroup under allele contrast, but failed to detect a significant association in the Caucasion population under all genetic models. This meta-analysis of the associations of the rs738409 polymorphism with NASH showed a significant relation. In the subgroup analysis stratified by ethnicity, similar correlations were observed in both Caucasian and Asian populations. The results from the allele contrast were consistent with those from the other genetic models. The sensitivity analysis revealed that no single study qualitatively changed the pooled odds ratios. These findings suggested that rs738409 polymorphism was strongly associated with NASH.In our meta-analysis, it appears that the rs738409 polymorphism is more likely to increase the NASH risk instead of simple steatosis. Consistent with our results, animal studies have revealed that, although PNPLA3 has triglyceride lipase activity and is responsible for the transalkylation of acylglycerol, knockout of PNPLA3 has no effect on liver steatosis or insulin resistanceDespite the inevitable limitations of this meta-analysis, we believe that our research provides useful information. First, the individual sample size of each study included in our meta-analysis was too small to obtain a definite association between rs738409 polymorphisms and NAFLD risk, but the pooled odds ratios generated from the 23 studies significantly increased the statistical power of the analysis compared to that obtained with a single study. Moreover, the protocol of this meta-analysis has been well-designed with explicit criteria and methods for study selection, data extraction and data analysis, which allowed reliable inferences about causality. Third, there was no significant publication bias in this meta-analysis, and the results of the sensitivity analysis support the stability of the results.However, some limitations of this meta-analysis should be addressed. First, the retrieved literature may not be sufficiently comprehensive. Only published case-control studies were included in this meta-analysis. Second, most of the study subjects were of Caucasian and Asian ancestry, and the Hispanic subgroup was very limited in this meta-analysis. Thus, potential selective bias and publication bias may have occurred. Third, because NAFLD was a multifactor disease, the potential effects of gene-gene and gene-environment interactions should be considered. Fourth, the sample size of NASH in this meta-analysis was so small that the statistical power for making a definitive conclusion regarding the possible risk of the rs738409 polymorphism was limited.In conclusion, results from this meta-analysis showed that the G allele at PNPLA3 gene was a risk factor for NAFLD and its subtype NASH, especially in Asian, Caucasian and Hispanic populations. However, no association was observed between the rs738409 polymorphism and simple steatosis risk. Further studies with higher quality, more participants and various ethnicities are needed to obtain a more precise estimate of the genetic effects.R.-F.X. and G.-Z.C. conceived the study design, and wrote the manuscript; A.-Y.T., S.-S.Z. and Y.-B.D. performed the analyses. All authors read and approved the final manuscript.Supplementary information


Tea and coffee are the most commonly consumed beverages in the worldwide. The relationship between tea and coffee consumption on the risk of laryngeal cancer was still unclear.Relevant studies were identified by searching electronic database (Medline and EMBASE) and reviewing the reference lists of relevant articles until Oct. 2013. Observational studies that reported RRs and 95% CIs for the link of tea and coffee consumption on the risk of laryngeal cancer were eligible. A meta-analysis was obtained to combine study-specific RRs with a random-effects model.A total of 2,803 cases and 503,234 controls in 10 independent studies were identified. The overall analysis of all 10 studies, including the case-control and cohort studies, found that tea drinking was not associated with laryngeal carcinoma . However, coffee consumption was significantly associated with the laryngeal carcinoma . A dose-response relationship between coffee intake and laryngeal carcinoma was detected; however, no evidence of dose-response link between tea consumption and laryngeal carcinoma risk was detected.The results from this meta-analysis of observational studies demonstrate that coffee consumption would increase the laryngeal cancer risk, while tea intake was not associated with risk of laryngeal carcinoma. It was reported that cancer continues to be a major public health problem despite of the development in medical technology for its prevention, diagnosis and treatment. Cancer of the larynx is fourteenth most common cancer in the world and it is the most common cancer in the head and neck Tea and coffee are the most commonly consumed beverages in the worldwide. Numerous studies have been conducted to investigate the association between their consumption and risk of kinds of cancers Meta-analysis is a useful statistical tool to pool the relevant studies together and gain a more powerful conclusion. The meta-analysis was also conducted for the searching for the potential causes of laryngeal cancer. For instance, Li X et al investigate the association between human papillomavirus (HPV) infection through combining the relevant studies together and they found that HPV infection, especially infection due to the high-risk type HPV-16, was significantly associated with the risk of laryngeal carcinoma We followed the Meta-Analysis of Observational Studies in Epidemiology (MOOSE) The studies were be considered included if they met the following inclusion criteria: 1) studies reported the association between tea or coffee or both and risk of laryngeal cancer; 2) studies obtained a case–control or cohort study design; 3) the value of relative risk (RR), odds ratio (OR) with 95% confidence intervals (CI) or the raw data to calculate them were reported The data extraction was conducted via a standardized data extraction form, collecting information on the name of first author, the publication year, study design, number of cases and controls, sample size, study site, gender of participants, type of the tea, tea drinking definitions, adjustments of the confounding factors, and the OR/RR value with 95% CI. When the tea drinking wasn't stated, it would be categorized as “ever versus never”. When the OR or RR was not reported in the article, the RR with 95% confidence intervals (CI) with the raw data and no confounding factors were adjusted. If only stratified results  were provided, fixed-effect methods were obtained to summarize the results into a single parameter for each study.The study quality was assessed by two reviewers back to back and any discrepancies were re solved by reevaluating the included articles and discussion with a third investigator. We obtained the Newcastle-Ottawa Scale (NOS) Assessment of the quality of the included studies. The study quality was assigned to each study based on the 3 parts: selection, comparability, and exposure and outcome condition. The NOS assessed the selection, comparability and exposure of a case-control study, while the selection, comparability and outcome of a cohort study. The study with more than 6 stars would be regarded in relative high quality.The RR was obtained to approximate RR in this meta-analysis because of the low incidence rate of laryngeal cancer. When both the crude and the adjusted OR/RR values were offered, only the adjusted value would be adopted for the meta-analysis. If only the raw data was reported, we would calculate the unadjusted RR.2 test and quantified with the I2 statistic. When P for the heterogeneity was <0.1 and 2I>50%, the interstudy heterogeneity would be considered statistically significant.The ORs and 95% CIs of all the included studies were pooled using the general variance-based method with a random-effects model. The heterogeneity among the included studies was measured by the χThe source of the statistically significant heterogeneity was assessed by both removing the included studies one by one to measure whether any single study was the source of the heterogeneity. Another independent method to detect the source of heterogeneity was to conduct a subgroup meta-analysis. Subgroup analyses were conducted by the study designs (case-control or cohort study), population of or hospital based design, prospective or retrospective study design, and the different sexual groups .We examined a potential linear dose-response relationship between drinks consumption (both tea and coffee consumption were considered) and risk of laryngeal carcinoma using restricted cubic splines with three knots at percentiles 25%, 50%, and 75%of the distribution. A two-stage random-effects dose-response meta-analysis tested was adopted for the detection of the potential nonlinear relation. A P value for nonlinearity was calculated by testing the null hypothesis that the coefficient of the second spline is equal to 0 A sensitivity analysis was performed by excluding the studies with a relative lower methodological quality. We would assess the effect of tea and coffee consumption and risk of laryngeal carcinoma through just including the high-quality studies. The publication bias was evaluated using funnel plots and the Egger test The flowchart of the study selection was presented in A total of 2,803 cases and 503,234 controls in 10 independent studies were identified. Among the 10 included studies, 3 reported the data of tea consumption, 3 reported coffee consumption, 3 reported both tea and coffee consumption and 1 pool the data of tea and coffee together. Overall, one study was a prospective population based cohort study and the rest 9 studies were retrospective hospital based case-control studies. The geographicical distribution of the studies sties was 6 were in European, 2 in Americas and 2 in Asia. The detailed age, gender distribution, categories of tea and consumption, and adjustments of confounding factors were demonstrated in Study quality was judged on the basis of the Newcastle-Ottawa Scale (1–9 stars). The scale distribution was from 5 to 8 stars. Among the 10 included studies, 8 studies demonstrated a relatively high quality (more than 6 stars in NOS) .The overall analysis of all 10 studies, including the case-control and cohort studies, found that tea drinking was not associated with laryngeal carcinoma . However, coffee consumption was significantly associated with the laryngeal carcinoma . When the relationship between tea and coffee consumption and risk of laryngeal carcinoma was considered, no significant association was detected in the meta-analysis  nor the subgroup analyses  .The subgroup analyses were conducted by the study designs, population or hospital based design, prospective or retrospective design, study sites and the tea drinking category. The effects of tea and coffee consumption and laryngeal carcinoma were detected discretely in the subgroup analyses. Tea consumption was not associated with laryngeal cancer in neither cohort  nor case-control studies . The similar results were detected in the subgroup analyses by the detailed study designs and the study site. When comparing the one who ever drank tea with never drank tea, tea consumption was unrelated with the laryngeal carcinoma . No association was detected when the moderate with the low ranks of tea drinking  .Among the 6 studies that reported the association between coffee consumption and laryngeal carcinoma, 1 study was a prospective population based cohort study, while the rest 5 studies were retrospective hospital based case-control studies. The subgroup analyses of the prospective population based cohort study demonstrated that coffee intake wasn't associated with the laryngeal carcinoma . However, this result should be considered with caution considering that only one study was included in this subgroup. The rest 5 study were pool as the retrospective, hospital based and case-control subgroups and a significant association was detected . In the subgroup analyses by the study site, a significant result was detected in the Europe subgroup  but not in the Americas . When the ever coffee drinker was compared with the never drinker, ever coffee drinking was associated with the incidence of laryngeal carcinoma . No statistically significant association was detected in the comparison between moderate and low coffee drinking .2 = 79.8%; P<0.001). The significant heterogeneity was detected when the association between tea  and coffee consumption  and the laryngeal carcinoma was tested. We tried to explore the source by excluding the included studies one by one and re-count the heterogeneity and the analyses did not identify any one study which contributed significantly to the heterogeneity. The advanced subgroup analyses by study designs, study sites, and case group definitions demonstrated no satisfactory results in the exploring the source of the significant heterogeneity. In advanced studies, the meta-regression showed no satisfactory results in this meta-analysis.The heterogeneity was statistically significant when all the studies were pooled together . A 1 pack of coffee consumption increment conferred a RR of 1.22 .When only the articles with a relative high quality (over 6 stars NOS stars) included in the meta-analysis, the results was notablely influenced. In the sensitivity analysis, the tea was not associated with laryngeal carcinoma , while coffee drinking increased the risk of laryngeal carcinoma . A significant heterogeneity should be noted as well (tea: IP = 0.352) and coffee consumption  and risk of laryngeal carcinoma.The funnel plot for both tea intake  and coffIn this current meta-analysis, a total of 2,803 cases and 503,234 controls in 10 independent studies were identified. The quantitative synthesis of these observational studies demonstrated that tea consumption was not associated the incidence of laryngeal carcinoma, while coffee consumption was associated the increased the risk of laryngeal cancer. When the association between tea consumption and risk of laryngeal cancer was considered, the subgroup analysis by the study designs, study sites and tea drinking category displayed no statistical significant results. However, when the association between tea consumption and risk of laryngeal cancer was considered, coffee consumption reduced the risk of laryngeal carcinoma in the retrospective hospital based case-control studies, while no association was detected in prospective population based cohort study. When the ever coffee drinker was compared with the never drinker, ever coffee drinking was associated with the incidence of laryngeal carcinoma. No statistically significant association was detected in the comparison between moderate and low coffee drinking. The heterogeneity was statistically significant when all the studies were pooled together, while no satisfactory results was demonstrated in the exploring the source of the significant heterogeneity. The results of the sensitivity analyses suggest that the conclusion was quite robust.The etiology of laryngeal cancer is quite unclear by now. In general, it is a disease caused by both genetic and environmental factors. By now, several genetically polymorphic enzymes like cytochrome P450 1A1 are reported to be related with laryngeal cancer Several previous meta-analyses have been conducted to explore the association between coffee consumption and cancer risk. Both a protective and a harmful effect of tea drinking were detected on the incidence of cancer were detected according to the cancer types Tea and coffee are served as hot beverage usually. They might modify the incidence of laryngeal cancer through more than the tea and coffee itself. In a systematic review, the results strongly suggest that high-temperature beverage drinking increases the risk of esophageal cancer There are no definite biological mechanisms of the potential harmful role of coffee on laryngeal cancer, thus there are still no explanations for the different effects of tea and coffee on laryngeal cancers. In particular, the coffee drinks contains several phenolic compounds , melanoidins and diterpenes, which might produce certain effects on the development of laryngeal cancer To our best knowledge, this is the first meta-analysis investigating the relationship between tea consumption and risk of laryngeal cancer. A comprehensive literature search and advanced detailed consulting the relevant references was conducted to make sure all the potential logical articles included. In this current meta-analysis, the results of the sensitivity analyses suggest that the conclusions of this study were quite robust. The stratified analyses and detailed dose-response analysis provided abundant knowledge.Despite these strengths mentioned above, some limitations of the current meta-analysis should be noted. Firstly, only 10 studies with relative low quality were included in this meta-analysis and the relative small sample size would make the conclusion unstable. We have tried our best to access all the possible studies. In the quantitative synthesis, the random-effect model might provide a more conservative conclusion. Secondly, the most of our studies followed a case–control study design, and therefore there were recall and selection bias which are inherent to retrospective studies. More relevant cohort studies are required in the future. Besides, considering the significance of between-study heterogeneity, the conclusions should be considered with caution.In conclusion, tea consumption wasn't associated with the risk of laryngeal cancer, while coffee consumption increased the laryngeal cancer risk. Nevertheless, because of the potential limitations of this meta-analysis, conclusions must be drawn with caution, and more well-designed studies with large sample sizes should be conducted for further validation.S1 ChecklistPRISMA checklist.(DOC)Click here for additional data file.


Some studies have suggested that vasectomy is associated with the increased risk of prostate cancer, however, this conclusion is not supported by all the published studies. In order to examine the relationship between vasectomy and prostate cancer risk, we conducted a meta-analysis of cohort studies to clarify this controversial association. PubMed and Medline were used to identify the cohort studies that reported the association of vasectomy with prostate cancer risk from 1980 to January 2015. Based on a random effects model, the RR and 95% CI were used to assess the combined risk. In total, 10 cohort studies involving more than 7027 cases and 429914 participants were included. There was no significant relationship between vasectomy and prostate cancer risk, the pooled RR (95%CI) was 1.11 (P = 0.109). In subgroup-analysis, the relationship between vasectomy and prostate cancer risk was not significantly modified by the length of follow-up and population distribution except Americans. Omission of any single study had little effect on the pooled risk estimate. Little evidence of publication bias was found. In conclusion, our meta-analysis suggests that vasectomy is not associated with the increased risk of prostate cancer. More studies based on other populations including the Chinese are needed. Vasectomy is a common and reliable way of permanent contraception among the adult men3It has seen a rapidly increasing interest in the relationship between vasectomy and the risk of prostate cancer in the past two decades. Many studies have examined this relationship567895769111315PubMed and Medline were used to identify the studies that reported the relationship between vasectomy and the risk of prostate cancer which were published from 1980 to January 2015. The following keywords were used: “vasectomy” in combination with “prostate cancer” and “prostate carcinoma”. The reference lists of the retrieved studies were reviewed. If necessary, we contacted the authors of the original studies for the required data.The studies that met the following criteria could be included: 1. the study had a cohort study design; 2. the study reported the association between vasectomy and the risk of prostate cancer; 3. the publication language was confined to English; 4. if there were duplicate publications on the same study population or authors, we included the most recent one. Firstly, we checked the titles and abstracts to confirm the potential studies. If uncertain, a subsequent full-text assessment was conducted. The study retrieval was conducted by two independent authors (Shang Y.G. and Han G.W.). All disputes were resolved by discussion.We conducted the data extraction with a standardized form. The following information was collected from the included studies: the last name of the first author, publication year, study population, number of cases and participants, length of follow-up, estimated effects from the most fully adjusted model and the corresponding 95% CI for the relationship between vasectomy and prostate cancer risk.Rather than reporting the aggregate scores, the key components of designs, such as selection of study populations, ascertainment of exposure and outcome, duration of follow-up, were used to estimated the quality of primary studies.2 statisticsAll statistical analyses were conducted with STATA version 12.0. The pooled RR and the corresponding 95%CI were calculated to assess the relationship between vasectomy and prostate cancer risk by using the assumptions of a random-effects model that considered within-study and between-study variation568131415161920The main characteristics of included 10 cohort studies are presented in 581314166658141516132 = 57.8%, P < 0.05).The RRs from the 10 cohort studies and pooled RR are presented in 58Meanwhile, we conducted a subgroup-analysis to identify the effect of population distribution on the pooled risk estimate. The results of this subgroup-analysis is presented in In addition, we assessed the effect of the length of follow-up on the pooled RR by conducting another subgroup-analysis . We did 2 = 0.1%, P = 0.428), which suggested that these 2 studies were the main source of heterogeneity. The result of sensitivity analysis is presented in By removing one study at a time, we conducted a sensitivity analysis to assess the influence of each included study on the pooled RR. The combined RRs were similar to each another, and none significantly modified the pooled RR. But after excluding 2 studies5The funnel plot showed all the included studies symmetrically distributed in the triangle area. Begg's and Egger's regression test showed a low probability of publication bias in our study (P = 0.878). The funnel plot of the studies is presented in Our meta-analysis suggests that vasectomy is not associated with the increased risk of prostate cancer. All included studies had a cohort study design, however, there was substantial evidence of heterogeneity, the different populations, varied characteristics of participants and length of follow-up might partially explain it. We searched the source of heterogeneity by conducting a sensitivity analysis and we found that 2 studies were the main source of heterogeneity5The subgroup-analysis showed that the relationship between vasectomy and prostate cancer risk was not significantly modified by the length of follow-up and population distribution except American males. This result was not wondrous, because 6 studies5813141658Giovannucci et al found a significant positive association of vasectomy with prostate cancer risk in American males8Vasectomy is a common method of male sterilization, whether it is linked to the increased risk of prostate cancer has been debated for many years. In 1980s, many case-control studies showed that vasectomy history was not linked to the increased risk of prostate cancer222425The present meta-analysis had some strengths. First, it was the updated one to assess the relationship between vasectomy and prostate cancer risk by combining cohort studies. We knew that in case-control studies, artificial association might appear due to inadequate selection of controlsMeanwhile, some limitations should be considered. First, residual confounders and unmeasured factors always consist in observational studies. Residual confounders or unmeasured factors are still possible after adjusting for the most relevant confounding factors6Our meta-analysis suggests that vasectomy is not associated with the increased risk of prostate cancer. More studies based on other populations including the Chinese are warranted.S.Y.G., C.D., H.G.W. and L.J. wrote and revised the main manuscript text; Z.J., L.C.C. and Y.S.H. prepared all the figures. All authors reviewed the manuscript.


P > 0.05). However, a meta-analysis among 11678 cases and 12861 controls showed that rs7529229-C allele was significantly associated with a decreased risk of CHD, especially in Europeans . Since there is significant difference among different populations, further studies are warranted to test the contribution of rs7529229 to CHD in other ethnic populations.The goal of our study is to test the association of IL6R rs7529229 polymorphism with CHD through a case-control study in Han Chinese population and a meta-analysis. Our result showed there is a lack of association between IL6R rs7529229 polymorphism and CHD on both genotype and allele levels in Han Chinese ( Coronary heart disease (CHD) is one of the leading causes of human deaths in the developed and developing countries such as China . As a co IL6R. IL6R signaling activates an intracellular signaling cascade leading to the inflammatory response [Interleukin-6 is an inflammatory cytokine , whose sresponse , 8. IL6R is located on 1q21, a susceptible locus for CHD. IL6R rs7529229 is a T/C variation associated with both IL6R level and a decreased risk of CHD events in Europeans [ IL6R rs7529229 polymorphism and CHD in Han Chinese. In addition, we performed a meta-analysis of the available case-control studies between rs7529229 of IL6R gene and CHD.Humann = 263) were diagnosed with the angiographic evidence that coronary artery stenosis was greater than 50% in one or more major coronary arteries [A total of 459 unrelated individuals were selected between May 2011 and November 2013 from Ningbo Lihuili Hospital, Zhejiang, China. Of these, 263 patients had CHD  and 196 patients were non-CHD controls . The patients had been examined by standardized coronary angiography according to the Seldinger method [arteries . Gensini IL6R gene. The sequences of the two allele-specific primers were 5′-GCGGCAGGGCGGCAATGTGGTCGTGGTGAGTTACCC-3′ and 5′-GATTACCGAATGTGGTCGTGGTGAGTTACCT-3′. The sequence of a reverse primer was 5'-TTTCTATGATTCCCTTTCACAGAGGTTTGA-3′. The reaction was performed with an initial denaturation stage at 95°C for 30 sec, followed by 40 cycles at 95°C for 30 sec, 59°C for 30 sec, and 72°C for 30 sec, and a final extension at 72°C for 30 sec. Genotyping of the PCR products was performed on the Roche LightCycler 480 Fluorescence Real-Time PCR System  using melting temperature shift (Tm-shift) according to the manufacturer's instructions [Human genomic DNA was prepared from peripheral blood samples using the nucleic acid extraction automatic analyzer  and was quantified using the PicoGreen dsDNA Quantification kit . Amplification was performed on the ABI GeneAmp PCR System 9700 Dual 96-Well Sample Block Module . Genomic DNA was subjected to polymerase chain reaction (PCR) with primers specific toructions , 13. Tm-ructions , 15 . The search keywords included “coronary heart disease” or “coronary artery disease” or “myocardial infarction” combined with “IL6R” or “interleukin-6 receptor” or “rs7529229” or “polymorphism” and “genetic association.” We read the full-text articles to collect the relevant information. References listed on the retrieved articles and previous meta-analyses on this subject were searched to appraise other studies of potential relevance. The included studies for the meta-analysis need to be case-control design and need to have information consisting of ORs and their 95% CIs or genotyping data to measure the relative risk. Data extraction was carried out by at least two reviewers (Xiaoliang Chen and Ping Peng) on a standard protocol, and the consensus data were established by discussion. In the meta-analyses, the following data collection was included: name of the first author, publication year, country, ethnic population, study stage, numbers of individuals in the case and the control groups, OR, and 95% CI.t-test was employed to analyze differences between two study groups. χ2 analysis was used to compare the categorical variables. Genotype and allele frequencies between CHD cases and healthy controls among the different subgroups were compared using χ2 test. Association between Gensini scores and rs7529229 was compared using linear regression test. Mann-Whitney test was used for the association of Gensini scores with rs7529229 under the dominant and the recessive models. The power of the study was estimated by the Power and Sample Size Calculation software (v3.0.43) [I2 statistic at the significant level of 0.05. The combined odds ratios (ORs) along with their 95% confidence intervals (CIs) were assessed with inverse-variance fixed-effect model  [P < 0.05 was considered to be statistically significant.Hardy-Weinberg equilibrium (HWE) was analyzed using the Arlequin software (v3.5) . The stav3.0.43) . Meta-anv3.0.43) . Heterog= 0.441) . SubgrouP < 0.05). However, there were no significant differences between the two groups for a series of biochemical parameters, including high-density lipoprotein cholesterol (HDL-C), triglycerides, low-density lipoprotein cholesterol (LDL-C), and total cholesterol.As shown in χ2 = 0.61, P = 0.73). The allelic distribution of rs7529229 did not differ between CHD cases and non-CHD controls . Our result suggested that rs7529229-C allele was associated with CHD risk. A future subgroup meta-analysis showed that rs7529229 of IL6R gene was a protective factor of CHD, especially in Europeans . Sensitivity analyses were repeatedly conducted when each particular study was omitted. As shown in P = 0.25) and Egger's regression plots .Our meta-analysis, including a total of 11,678 cases and 12,861 controls, examined the association between the rs7529229 polymorphism and CHD risk. We found that rs7529229 of IL6R was investigated in the present study. According to the report in dbSNP, there were at least 256 SNPs on the IL6R gene locus. Therefore, the results of IL6R rs7529229 might not stand for the rest of the IL6R SNPs. Thirdly, we only searched the literatures in Chinese for the eligible research included in the meta-analysis. Meanwhile, case-control studies with negative results were more likely to be unpublished. Potential language and publication bias might exist in the meta-analysis.There are several limitations in our study. Firstly, the power of our case-control study only reached 12.2% at alpha level of 0.05, so we could not exclude the possibility of lack of power in our study mainly due to the relatively small sample size. Secondly, only one polymorphism of IL6R with the risk of CHD. Further investigation on other SNPs on the gene is warranted to validate our findings in the Chinese population.In conclusion, our meta-analysis has established a strong contribution of rs7529229-C allele to reduced risk of CHD, especially in Europeans, although our case-control study is unable to find association of the


SIRT1 exerts protective effects against endothelial cells dysfunction, inflammation and atherosclerosis, indicating an important role on myocardial infarction (MI) pathogenesis. Nonetheless, the effects of SIRT1 variants on MI risk remain poorly understood. Here we aimed to investigate the influence of SIRT1 polymorphisms on individual susceptibility to MI. Genotyping of three tagSNPs  in SIRT1 gene was performed in a Chinese Han population, consisting of 287 MI cases and 654 control subjects. In a logistic regression analysis, we found that G allele of rs7069102 had increased MI risk with odds ratio (OR) of 1.57  after adjustment for conventional risk factors compared to C allele. Similarly, the combined CG/GG genotypes was associated with the increased MI risk  compared to the CC genotype. Further stratified analysis revealed a more significant association with MI risk among younger subjects (≤ 55 years old). Consistent with these results, the haplotype rs7069102G-rs3818292A-rs4746720T containing the rs7069102 G allele was also associated with the increased MI risk . However, we did not detect any association of rs3818292 and rs4746720 with MI risk. Our study provides the first evidence that the tagSNP rs7069102 and haplotype rs7069102G-rs3818292A-rs4746720T in SIRT1 gene confer susceptibility to MI in the Chinese Han population. Myocardial infarction (MI) is the world’s leading cause of morbidity and mortality, with the World Bank estimating that the number of individuals with MI in China will increase to 23 million by 2030 ,2. Lots Sirtuin1 (SIRT1), the closest homolog of yeast silent information regulator 2 (Sir2), functions as a NAD+-dependent histone deacetylase  after adjustment for conventional risk factors compared to C allele. Similarly, the combined CG/GG genotypes was associated with the increased MI risk  compared to the CC genotype. Taken together, our data indicated that SIRT1 tagSNP rs7069102 may be associated with MI risk, and that individuals carrying G allele may have significantly increased MI susceptibility. However, we did not detect any association between rs3818292 or rs4746720 and the risk of MI in allelic or genotypic analyses  carrying G allele  in 287 MI patients and 654 controls. Our result showed that SIRT1 rs7069102 G allele is associated with a significantly increased risk of MI. We did not detect any association between rs3818292 and the risk of MI in allelic or genotypic analyses, which was in line with the previous report of rs3740051 , effectively enabling us to reduce the number of markers needed to analyze the whole gene. Thus, we should keep in mind that the association of the rs7069102 polymorphism with the risk of MI may be due to a direct causative effect of this SNP, or because it is in LD with other functional variants located in or near the SIRT1 gene or its regulatory regions. Future fine mapping of the SIRT1 gene may detect such functional variants. Finally, given that the results of the present study were not replicated, further studies in different population could help to validated the significance of the association between these tagSNPs and the risk of MI. However, our observations provided valuable insights and interesting information and might serve to guide future studies in this area.Several limitations need to be addressed in this case-control study. First, the patients and controls were enrolled from hospitals and may not represent the general population. Nonetheless, the genotype distribution of the controls was in Hardy-Weinberg equilibrium. Second, although the statistical power  of rs7069102 is not very low, more samples should be recruited to improve the statistical power of our analysis. Third, the tagSNPs we selected are non-coding, and we therefore assume these variants to be linked with one or more functional variants within the SIRT1 tagSNP rs7069102 and the haplotype (rs7069102G-rs3818292A-rs4746720T) are associated with an increased risk of MI. Our finding suggests that the genetic polymorphism within SIRT1 gene may play a role in the occurrence of MI in a Chinese population, although further studies with larger sample size and in diverse ethnic populations are required to confirm the general validity of our findings.In summary, our study provides the first evidence that the G allele of S1 FigSIRT1 gene structure. SIRT1 gene is composed of 9 exons and spans 33.72kb. The exons are represented as dark gray boxes. D’ values are plotted as a graph to show linkage disequilibrium between the three tagSNPs in S1B Fig. Details of the picked tagSNPs and respective alleles captured are also provided in S1B Fig.S1A Fig. provides the details of the (TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.S1 Table(DOC)Click here for additional data file.S2 Table(DOC)Click here for additional data file.S3 Table(DOC)Click here for additional data file.