Certain Coffee is one of the most popular and consumable drinks worldwide. However, there are conflicting results on the influence of this drink in oral and pharyngeal cancer risk. To clarify this, we aimed to systemically review and carry out a meta-analysis of the relevant literature on the association between coffee and oral and pharyngeal cancer.We carried out an electronic search of publications up to August 2016 from PubMed, National Library of Medicines Medline, Embase, Science Direct and the Cochrane Central Register. The Newcastle–Ottawa scale was used to address the quality of the studies a meta-analysis was carried out using random-effects models.p<001) for high versus low coffee consumption with a moderate heterogeneity . Regarding studies on oral cavity cancers we observed a pooled OR of 0.82; 95% CI =.58-1.16; p=.257) and for pharyngeal cancers a pooled OR of .72 . There was no significant publication bias.From the 22,515 entries identified in the search, 13 case-control and 4 cohort studies were selected. With regards to quality on the Newcastle-Ottawa scale, an overall value of 6.06 was obtained. The analysis for oral and pharyngeal cancer grouped together indicated a pooled OR of .69 (95% CI of .57-.84; The results show an inverse association between high coffee consumption and the risk of oral and pharyngeal cancers, which indicates that coffee may have a protective role against these cancers. Further larger prospective observational cohort studies are needed to address any effect of other possible co-factors. Key words:Coffee, caffeine, oral cancer, pharyngeal cancer, risk, meta-analysis. Oral and pharyngeal cancer are the sixth most common cancers worldwide with a high mortality rate and overall poor for the survivors ,2. The aCoffee, after water, is the leading beverage in the world which highlights the importance of knowledge of its possible influence on human health . The comThe International Agency for Research On Cancer (IARC) in 1991 had previously classified coffee as possibly carcinogenic to humans (Group 2B). The large body of evidence currently available led the IARC to recently re-evaluate the carcinogenicity of coffee drinking. They concluded that there was inadequate evidence for the carcinogenicity of coffee drinking but did not refer to any protective effects of coffee consumption on oral or pharyngeal cancers .The common use worldwide of this beverage led to several studies that tried to analyse the contribution of coffee consumption on the risk of oral cancer, although with conflicting results. Some studies reported an association of high coffee consumption to an augmented risk of oral cancer ,9 and otTo answer this question, we aim to assess the impact of coffee drinking on oral cavity and pharyngeal cancer, by reviewing the relevant literature and carrying out a meta-analysis.A systematic review and meta-analysis was carried out to retrieve and summarise the published literature on coffee consumption on the development of oral and pharyngeal cancer. An initial electronic search of the databases was conducted up to August of 2016, within PubMed, Embase, Science Direct and the Cochrane Central Register. We used a combination of keywords and control terms (MeSH) were used wherever possible. The search terms included “coffee” OR “caffeine” OR “beverage” OR “drink” OR “lifestyle” AND “mouth cancer” OR “oral cancer” OR “pharyngeal cancer” OR “pharynx cancer” OR “oropharynx cancer” OR “nasopharynx cancer” OR “hypopharynx cancer” OR “oropharyngeal cancer” OR “nasopharyngeal cancer” OR “hypopharyngeal cancer” AND “risk”. Additionally, other relevant sources were screened, including the reference list from the pool of included studies and on relevant review papers on the topic.The inclusion criteria included observational studies  in English, Portuguese, French and Spanish languages with regards to squamous cell carcinoma involving the oral cavity or pharynx and the association with coffee consumption. In order to eliminate the language bias, if an abstract or article was written in a different language, a request for translation of the article into English was carried out before allowing for a further assessment. Information on odds ratio (OR) or relative risk (RR) and respective 95% confidence interval (CI) or information to calculate them were considered mandatory. All articles which did not fit into the criteria were excluded, including reviews, author debates, letters to the editor or studies in which it was not clear of the impact of coffee consumption in the development of oral or pharyngeal cancer.The title and abstracts from each article were identified via the search engines and were independently reviewed by the authors  for consideration of inclusion within the systematic review. If insufficient information was provided in the abstract or if there was any doubt to include an article, then the full text was reviewed before reaching a consensus.- Data ExtractionIndividual measures were extracted from each selected study, using a standardised form. For each article, the following informa-tion was extracted: the principal author’s last name, year of publication, experimental design, country, sample size, cancer type, OR´s or RR´s with respective 95% CI´s regarding coffee consumption (highest vs lowest consumption) and covariates. Three main areas of analysis were performed: the first with the goal of estimating the overall effect of coffee consumption on oral cavity and pharyngeal cancers together, the second focused on the relationship between coffee consumption and oral cavity cancer , and the last on the association between coffee consumption and pharyngeal cancer .- Quality assessmentWe evaluated the methodological quality of the selected studies by using the Newcastle–Ottawa scale (NOS) for observational studies. The check-list contained 9 items  with every item accounting for 1 point. Articles were independently assessed by authors  and in case of disagreement, a final consensus would be reached. Supported by previous studies , article- Statistical analysisThe meta-analysis procedures were carried out by an experienced independent statistician. For each study, the meta-analytic procedures were implemented with Stata® software . The Odds-Ratios (ORs) and the corresponding 95% CI were extracted from individual studies. For studies in which the effects were presented separately for different subgroups , weighted averages of estimates were calculated. The presence of significant heterogeneity was statistically tested with the Cochran Q-test and I2 statistics. I2 values of 25, 50 and 75 represent low, moderate and high heterogeneity, respectively. In the presence of heterogeneity, subgroup analyses were conducted to explore whether the study design, country and number of cases had an impact in heterogeneity levels.For each individual study, the OR for the highest versus the lowest level of coffee consumption were extracted. Using this approach, an OR value of 1 indicates an equal risk of oral and pharyngeal cancer between highest and lowest categories of coffee consumption; an OR greater than 1 demonstrates an increased risk of oral and pharyngeal cancer in individuals consuming the highest quantities of coffee; an OR lower than 1 indicates a decreased risk of oral and pharyngeal cancer in individuals consuming the highest quantities of coffee.p<0.05 represents statistically significant publication bias). The estimation of possible missed studies was statistically tested with the Duval and Tweedie non-parametric “trim and fill” method.DerSimonianand Laird random-effects models were used to determine the overall estimates. The presence of publication bias was examined through the visual inspection of funnel plot asymmetry, and statistically tested using the Begg and Mazumdar rank correlation method , with the odds for cancer risk in individuals with the highest coffee consumptions are 1.45 times less than the odds for cancer risk in individuals with the lowest coffee consumptions . The results are graphically represented in the forest plot . Additionally, it was noted that the overall effect remains statistically significant . For the group of cohort studies, the level of heterogeneity was slightly higher (I2=54.7%) than the group of case-control studies, and non-significant (Χ2 (3) =6.62, p=.085). However, in contrast with the group of case-control studies, it was observed that the level of coffee consumption was not related with the presence/absence of oral/pharynx cancers .In order to further explore the heterogeneity, subgroup analyses were conducted. When considering case-control studies alone, it was observed that the heterogeneity had a small decrease comparing with the overall analysis  =22.42, p=.049). On the other hand, the level of heterogeneity was less pronounced in Asian  =5.69, p=.128) and European  =13.00, p=.072) countries. The summary of the other results of the subgroup analyses is presented in The conduction of meta-analyses stratified by the location of the country revealed a significant heterogeneity between studies conducted in the Americas  =9.56, p=.159), whereas for studies with 300 or less cases, the heterogeneity between studies increased to moderate, and significant, levels  =22.96, p=.006).With respect to the studies’ sample size, it was evident that studies with more than 300 cases had lower (I2=35.3%) and non-significant heterogeneity levels (Χ2 (6) =9.27, In order to ascertain the impact of each study on the overall results, a sensitivity analyses were performed with a leave-one-out strategy . However, it was observed that the exclusion of a single study did not produce significant differences in overall effects when analysing the samples of oral and pharyngeal cancer together.To explore a possible specific relationship of the coffee consumption with cancer location, in particular for oral cavity vs the pharynx, we performed a subgroup analysis regarding the oral cavity vs pharynx locations separately.p=.257). The results are graphically represented in Figure p=.056) and moderate (I2=53.6%) heterogeneity was found across these studies.Among oral cavity cancer studies, the analysis with fixed effects revealed an absence of a statistically significant association between the consumption of higher levels of coffee and oral cancer risk . Furthermore, the level of hete-rogeneity among studies is reduced to a lower, non-significant level (Χ2(4)=5.46 p=.244, I2=26.7%), enabling the use of a fixed-effects model. Despite yielding a lower impact, it was also verified that the exclusion of the study of Bundgaard et al.  revealed that the exclusion of the study by Franco et al.  has a sid et al.  also conp=.019). The results are graphically displayed on the forest plot  .Among the studies analysing pharyngeal cancers alone, we observed a significant association between coffee consumption and risk for cancer  and Egger’s test (t(16)=-.1.15, p=.267) revealed that there was no significant publication bias. Furthermore, using the “trim and fill” method, no missing studies were identified.Begg’s funnel plot was used to assess the publication bias for the included articles. The graphical representation suggests a re-duced asymmetry Fig. . Begg’s Beverages such as alcohol and mate drinking have been linked to an increased risk of oral and pharyngeal cancers . AccordiThe chemical and biological properties of coffee could explain in part the positive influence on oral and pharyngeal cancer risk reduction. Coffee beverages, additionally to caffeine, contains a variety of antioxidant and anti-mutagen agents including phenolic derivates (such as chlorogenic acid and polyphenol caffeic acid) and diterpenes (such as cafestol and kahweol), that could act as carcinogenic detoxifying agents on oral and pharyngeal mucosa . FurtherThere are some limitations which have arisen during the execution of our meta-analysis. There was little information related to the type of coffee beans (Arabica or Robusta) or the brewing procedure, methodology used in the different studies we consulted, as well as the concentration of caffeine, and the size of the cups used. Moreover, there were different categories of coffee consumption in the studies, which did not allow for quantification of the association between oral cancer and the number of cups of coffee consumed. This meant that we only extracted the OR value to thereby compare the highest rate of consumption with the lowest rate. Another important aspect is the temperature of the coffee that almost in every study is not a controlled aspect. The temperature of oral beverages has been linked to some types of cancer ,20. ThisNevertheless, we think that our results are of interest and important for the comprehension of the role of coffee consumption on oral / pharyngeal cancers. First we have selected the case-control and cohort studies with a high quality score and large samples of cases and participants and performed an analysis reporting oral cavity and pharyngeal cancers but also analysing into a subgroup of oral cavity vs pharyngeal cancers. Secondly, we found no publication bias in our study.We can conclude that the results obtained from this meta-analysis, confirm that there is an inverse association between high coffee consumption and the risk of oral/pharyngeal cancer, so we can infer that individuals who consume higher amounts of coffee are less likely to develop oral cancer. However, in the future, it may be necessary to determine the robustness of the results obtained in this study and to thus confirm the inverse association demonstrated, as well as establishing greater rigour and consistency between variables and co-variables, especially tobacco and alcohol.


C4, C4A, and C4B gene copy number variations (CNVs) and susceptibility to autoimmune diseases, the results remain inconsistency for those diseases. Thus, in this study, a comprehensive meta-analysis was conducted to assess the role of C4, C4A, and C4B CNVs in autoimmune diseases in different ethnic groups. A total of 16 case-control studies described in 12 articles (8663 cases and 11099 controls) were included in this study. The pooled analyses showed that a low C4 gene copy number (GCN) (<4) was treated as a significant risk factor  for autoimmune diseases compared with a higher GCN (>4). The pooled statistical results revealed that low C4 (<4) and low C4A (<2) GCNs could be risk factors for systemic lupus erythematosus (SLE) in Caucasian populations. Additionally, the correlation between C4B CNVs and all type of autoimmune diseases could not be confirmed by the current meta-analysis . These data suggest that deficiency or absence of C4 and C4A CNVs may cause susceptibility to SLE.Although several studies have investigated the association between  C4 plays a pivotal role in the activation of immune defenses and the clearance of immune complexes or apoptotic debris in vitro and in vivo2C4 gene occurs, it could cause an aberrant immune response and an autoimmune or inflammatory disorderC4 gene is mapped in the class III region of the major histocompatibility complex (MHC) on chromosome 6p21.3 and there are two isotypes of C4 gene products, which are acidic C4A and basis C4BC4A and C4B isotypes can be distinguished by four specific amino acids at positions 1101–1106 (or 1120–1125 if the protein numbering starts at the initiation codon for Methionine), which were the results of five nucleotide polymorphisms6C4A prefers to bind to amino groups of immune complexes, while C4B binds to hydroxyl groups when activated8C4, C4A, and C4B CNVs could affect the strength of an individual’s immunity and susceptibility to autoimmune diseases81011The complement system, which is involved in both innate and adaptive immunity, is characterized by triggered-enzyme cascades activated by alternative, lectin or classical pathwaysC4 deficiency is one of the most frequent immunoprotein deficiencies in Homo sapiensC4A and C4B have various GCNs according to the Database of Genomic Variants (http://projects.tcag.ca/variation). The C4 gene encodes for either C4A or C4B, so the GCN of total C4 is equal to the sum of the GCNs of C4A and C4BC4 genes. C4 exists as two isoforms encoded by different genes, C4A and C4B. Together with three neighboring genes, C4 forms a genetic unit called the RCCX module (RP-C4A-CYP21-TNX or RP-C4B-CYP21-TNX). In the human diploid genome, the total C4 GCN varies from 2 to 6, with 4 being the most common. C4A GCN varies from 0 to 5, while that of C4B varies from 0 to 4, with 2 being the most common for both genesC4 GCNs, 20% have five or six, and <2% have only two8C4 GCNs, and more than 63% have two C4A and C4B GCNs18C4, C4A and C4B in all racial groups are four, two, and two, respectively.CNVs in the human genome belong to a significant sequence variation, indicating that large stretches of DNA may exist in a number of different forms in different individualsC4 are strongly associated with increased risk of developing SLEC4, C4A, and C4B CNVs in SLE and other systemic autoimmune diseases and determined whether C4 genes are a genetic master key to autoimmunity. Subgroup analyses focusing on disease type and ethnicity were conducted to investigate potential sources of heterogeneity in the included studies.Deficiencies of The literature selection process is shown in 1718192122232425262728The potential bias assessment of the included studies is presented in C4, C4A, and C4B CNVs and autoimmune diseases was investigated, and the detailed results of pooled odds ratios (ORs) and a stratification analysis are presented in C4: I2 = 85.3%, PH < 0.001; C4A: I2 = 89.5%, PH < 0.001; C4B: I2 = 72.8%, PH < 0.001). Hence, the meta-analysis detected a random-effect model. As presented in C4 GCN and autoimmune diseases. Individuals with low C4 copy numbers (<4) were more likely to develop an autoimmune disease . In the subgroup analysis involving disease type, the risk of SLE was clearly increased in individuals with low C4 GCNs . Moreover, the subgroup analysis involving ethnicity revealed that low C4 GCNs were significantly associated with the risk of autoimmune diseases among Caucasian individuals , while such an association was not confirmed in an Asian population . Furthermore, the results suggested that low C4A GCNs (<2) are more associated with apparent risk for autoimmune diseases compared with higher GCNs (≥2)  . No significant association between low C4B GCNs (<2) and autoimmune diseases was found in this meta-analysis  (The relationship between 10–1.94) . The sub93–1.24) .et al.’s reportC4 or C4A GCNs . Finally, the studies were stratified by ethnicity, which found an apparently increased risk of autoimmune diseases associated with low C4 or C4A GCNs in Caucasians . These results suggest that, regardless of the random effect model or fixed effect model used for meta-analysis, the pooled data was consistent, believable, and stable.The fixed effect model was applied to these data in accordance with Chen As suggested in C4, C4A, and C4B CNVs. As presented in C4, we performed a funnel plot using the trim and fill method. LnOR and 95% CI were 0.376  and 0.246 , respectively, before and after applying the trim and fill method, which indicated that publication bias was present in the meta-analysis  or C4A (<2) are predisposed to autoimmune disorders in the presence of environmental triggers. Consistent with previous studies, both C4 and C4A CNVs were regarded as pivotal genetic factors in the pathogenesis of SLE. Furthermore, our meta-analysis demonstrated that low GCNs of C4 (<4) or C4A (<2) could lead to increased risk of autoimmune diseases among Caucasian populations. In other words, differences in the association between different ethnicities may result from other factors, such as geography, socioeconomic development, or race.In our meta-analysis of 8663 cases and 11099 controls in 16 studies from 12 articles, we drew a general conclusion that C4 plays an essential role in innate and adaptive immune responses, which are involved in the classical and mannose-binding lectin complement activation pathways and help to direct against external attacks such as microbial infection, clearance of immune complexes, and removal of apoptotic cells31C4 protein is encoded by two polymorphic genes, C4A and C4B, which are located in the MHC on chromosome 6. C4 deficiency and dysfunction are linked to the pathogenesis of many autoimmune and inflammatory diseasesC4 deficiency may be one of the most penetrant genetic risk factors of SLE421Complement C4A and C4B are two isotypes of C4. Although they share >99% of their amino acid sequences, their chemical reactions to substrates are remarkably different. C4A tends to combine amino group-containing antigens or immune complexes, while C4B combines hydroxyl group-containing antigens. Because C4A more efficiently handles immune complexes, deficiency of C4A while C4B affects the development of SLE34et al.et al.C4 GCN and serum C4 protein concentration as well as C4A and C4B gene dosage and serum C4A and C4B concentration. In their study, a low GCN group  had a significantly lower serum protein concentration than that of a medium GCN group  and a high GCN group C4 plays a pivotal role in innate and adaptive immune responses, individuals with low GCNs of C4 (<4) or C4A (<2) could show reduced serum C4 or C4A concentrations, resulting in dysfunction in the ability to resist microbial infection, clear immune complexes, and remove apoptotic cells. Thus, low C4 CNV is a risk factor for SLE. As we know, SLE, RA, BD, and AS are immune-mediated autoimmune diseases. In this meta-analysis, we planned to summarize the current knowledge related to the relationship between C4 CNVs and SLE and other systemic autoimmune diseases and to determine whether C4 is a genetic master key to autoimmunity. Liu et al.C4 and those with less than 2 copies of C4A and C4B tended to be at less risk for GD. Liu et al.C4A may be associated with high risk for vitiligo in patients with GD. Hou et al.C4 GCNs (>4) and decreased expression of serum C4 in patients. Hou et al.C4A lead to risk for BD but not acute anterior uveitis by modulating the expression of C4A and enhancing IL-6 production. In another study, Lintner et al.C4A and C4 GCNs  among JDM patients and controls, suggesting that complement C4A deficiency appeared to be an important risk factor for JDM. Furthermore, Rigby et al.C4B deficiency resulted in increased risk for RA and had broad implications for the pathogenesis of RA. However, data from pooled analyses of C4 and non-SLE autoimmune diseases are relatively insufficient. Additionally, data from several published articles were only done by single method without validation using the other methods and therefore further studies with different methods and larger samples including individuals from different ethnic backgrounds would help reveal the role of C4 in GD, uveitis, JDM, and RA.It is generally accepted that CNVs are a genetic determinant of phenotypic variation8C4, C4A, and C4B CNVs and autoimmune diseases need to be addressed. First, heterogeneity among ethnic groups or disease types was discovered when the association between C4, C4A, and C4B CNVs and autoimmune diseases was investigated. However, based on the results of the sensitivity analysis, it is clear that the overall results were not affected by heterogeneity. Second, an obvious publication bias was detected in the comparison between C4 CNV and autoimmune diseases. Third, the size of the patient and control groups was relatively small in each study; therefore, future studies should employ a much larger sample size and include individuals from different ethnic populations. Fourth, the effects of common confounding factors, including sex, age, and medical condition, were not assessed in the present study because of insufficient data. Fifth, autoimmune diseases at various sites may vary widely in terms of the etiology of pathology, including genetic factors, immunologic factors, and environmental factors. Finally, the electronic databases from which we selected eligible studies were listed in English and Chinese; thus, the meta-analysis may have a language bias.Some limitations of the current meta-analysis of the potential relationship between C4 or C4A CNVs and susceptibility to autoimmune diseases, especially for Caucasian individuals with SLE.In conclusion, the present meta-analysis provides evidence-based pooled data revealing a significant association between low The Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) criteria were using as a guide for this meta-analysisC4, C4A, or C4B in relation to autoimmune diseases; (2) a case-control study design; (3) presence of enough data to calculate an OR and corresponding 95% CI; and (4) inclusion of patients diagnosed according to international standards. The following items were regarded as the primary exclusion criteria: (1) focus on families or twins rather than random general populations; (2) inclusion of individuals with overlapping investigated diseases; (3) a case study design; (4) insufficient data applied after emailing the relevant corresponding author. When several articles with the same types of patients were identified, the most detailed article with the largest sample size and most comprehensive analysis was included in the meta-analysis.Studies selected from electronic databases were included when they met the following criteria: (1) focus on the CNVs of total Data from the retrieved studies were extracted independently by two authors (N.L. and J.Z.). The following information was collected from each eligible study: first author, year of publication, country of origin, ethnicity of subjects, type of disease, sample size, genotyping methods, and information about GCN distribution in cases and controls. Two authors carefully screened the collected data and reached agreement in all respects. When the authors disagreed, a third reviewer (D.L.) weighed the arguments and then helped reach a consensus.Evaluation of the quality of extracted studies was also performed by two authors (N.L. and J.Z.) based on the HuGENet HandbookC4, C4A, and C4B CNVs and susceptibility to autoimmune diseases. The diversity of genotypes  among cases and healthy controls was compared. The heterogeneity between studies was estimated and measured by Cochran’s Q statistic as well as the I2 statistic. When the p value of chi-square statistic was less than 0.05 or the I2 value was more than 50%, the random-effect model was adopted for meta-analysis, indicating that heterogeneity existed across studieset al.ORs and the corresponding 95% CIs were estimated to evaluate the amount of correlation between How to cite this article: Li, N. et al. Association between C4, C4A, and C4B copy number variations and susceptibility to autoimmune diseases: a meta-analysis. Sci. Rep.7, 42628; doi: 10.1038/srep42628 (2017).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


DNMTs  were associated with occurrence or decrease of various tumors. However, the associations between DNMTs variations and gastric cancer (GC) risk were still conflicting. We aimed to assess the effect of DNMTs polymorphisms on the susceptibility to GC.Increasing studies showed that abnormal changes in single nucleotide polymorphisms (SNPs) of DNMT1, rs1550117, rs13420827 in DNMT3A, rs1569686, rs2424913 in DNMT3B). Four genetic models  were used. Moreover, a meta-sensitivity and subgroup analysis was performed to clarify heterogeneity source. Lastly, 17 SNPs that couldn't be meta-analyzed were presented in a systematic review.Firstly, we did a meta-analysis for 7 SNPs  for 7 SNPs showed that GC risk increased in rs16999593  and rs1550117  but decreased in rs1569686 . The remaining SNPs were not found associated with GC risk. Furthermore, the subgroup analysis indicated that for rs1550117 and rs1569686, the significant associations were particularly found in people from Chinese Jiangsu province  and that PCR-RFLP was a sensitive method to discover significant associations . Lastly, a systematic review on 7 studies for 17 SNPs suggested that rs36012910, rs7560488 and rs6087990 might have a potential effect on GC initiation.This meta-analysis demonstrated that rs16999593 and rs1550117 could contribute to GC risk and that rs1569686 might be a protective factor against gastric carcinogenesis. By using these SNPs as biomarkers, it is feasible to estimate the risk of acquiring GC and thus formulate timely preventive strategy. •DNMT1, DNMT3A and DNMT3B polymorphisms are associated with gastric cancer (GC) risk, in which rs16999593 and rs1550117 might contribute to GC and rs1569686 could be a protective factor against carcinogenesis.•DNMTs polymorphisms were particularly found in people from Chinese Jiangsu province.Significant •DNMTs polymorphisms.PCR-RFLP was a sensitive method to discover significant Research in context Genetic factors play a crucial role in GC risk. DNMTs genes were associated with occurrence or decrease of various tumors. But the effects of DNMTs on gastric cancer (GC) were not clear. Now our results proved that two and one variations in DNMTs were associated with GC, indicating a range of effects from the increased to the reduced. By using these variations as biomarkers, it is feasible to estimate the risk of acquiring GC and thus formulate timely preventive strategy. GC is arcinogen , only a rcinogen .DNMT1 gene which locates on chromosome 19p13.2 , which could cause DNA hypo-methylation or hyper-methylation  of hylation , are corhylation  such as hylation . Howeverflicting . Therefo22.1We did a literature search of PubMed, MEDLINE, Embase, Sinomed, CNKI, and WanFang databases to identify relevant studies up to June 1, 2016, using the search strategy: (stomach OR gastric) AND (neoplasms OR tumors OR cancers OR carcinomas) AND (DNMT1 OR DNMT3A OR DNMT3B OR DNMTs OR DNA methyltransferases). The languages were limited to English and Chinese. The search strategy for PubMed was listed in 2.2DNMTs polymorphisms and GC risk; (b). case-control or cohort studies. In addition, exclusion criteria were as follows: (a). reviews or meta-analysis; (b). overlapped articles or studies with overlapping data.All studies included in the meta-analysis were accorded with the following inclusion criteria: (a). study focused on the association of 2.3DNMT genes, SNPs, number of cases and controls, and value of HWE. To ensure accuracy of the data, inconsistencies were discussed with another reviewer until reach a consensus.Two investigators independently extracted the following data: first author, year of publication, province/country of origin, ascertainment of cases, source of controls, genotyping methods, 2.4The quality of each study was assessed according to the quality assessment criteria (Table S1) , in whic2.5DNMTs polymorphisms and the GC risk was evaluated by pooled odds ratios (OR), 95% confidence interval (95% CI) and P value of Z test (POR). If 95%IC across 1 or POR < 0.05, a significant association existed. Then if OR or 95%IC < 1, the mutant gene was a protective factor; otherwise, it was a risk factor. Heterogeneity was analyzed using the P value of Q test (Phet) and I2. If Phet < 0.1 or I2 > 50%, a significant heterogeneity existed. And then a sensitivity analysis and a subgroup analysis were performed. Sensitivity analysis was conducted through omitting one study by turns  was used to perform all analysis. We used four types of genetic models : homozygby turns , if the 33.1DNMTs SNPs and GC risk. Then 25 full-text articles were obtained to be assessed, in which 5 articles were excluded because 1 was duplicate publication and 4 did not contain information on DNMTs SNPs and GC risk. Ultimately, 20 eligible studies , so 13  studies . Among t studies . The mai3.2DNMTs polymorphisms and gastric carcinogenesis were shown in DNMT1 and DNMT3A, GC risk increased. For rs16999593, there was an association under heterozygote and dominant models  but not homozygote and recessive models . For rs1550117, the increased GC risk was discovered under homozygote, dominant and recessive models  but not heterozygote model . Conversely, GC risk decreased in DNMT3B. For rs1569686, the association was found under dominant model  but not heterozygote, homozygote and recessive models . Except all of the above, for rs2228611, rs8101866, rs13420827 and rs2424913, no significant associations were observed among all of the genetic models. Lastly, for SNPs not able to be quantitatively synthesized, the systematic review presented their associations with GC  and rs1569686 . A sensitivity analysis was conducted to explore which study primarily influenced the pooled ORs . For rs1550117, the heterogeneity was mostly caused by a study . Likewise, for rs1569686, I2 0%, Phet 0.88). We compared characteristics of the two studies to the other's. Two factors were screened out to explain the heterogeneity: population areas (Jiangsu province or others) and genotyping methods (PCR-RFLP or others). Then a subgroup analysis was performed  to the reduced (DNMT3B).Of the seven SNPs, two (rs16999593 and rs1550117) and one (rs1569686) were significantly associated with GC risk indicating a range of effects from the increased  and Key Specialty Foundation of The State Administration of Traditional Chinese Medicine (NO:ZJ0901ZL020). The sponsor had no role in study design, data collection, data analysis, data interpretation, or writing of the report.This work was supported by The authors have no conflicts of interest.


AA: odds ratio [OR] = 1.41, 95% confidence interval [CI] = 0.75-2.26; AG vs. AA: OR = 1.13, 95% CI = 0.74-1.73; AG/GG vs. AA: OR = 1.17, 95% CI = 0.75-1.81; GG vs. AA/AG: OR = 1.38, 95% CI = 0.79-2.42). Similarly, in the subgroup analyses by cancer type, ethnicity, and smoking status, no significant association with any genetic model was observed. In conclusion, the results of our meta-analysis suggest that the GSTP1 Ile105Val polymorphism is not associated with the development of gynecological cancer.The association between the glutathione S-transferase P1 (GSTP1) Ile105Val polymorphism and gynecological cancer susceptibility has been evaluated in many studies. However, the results remain controversial. Thus, this meta-analysis, based on 10 published case-control studies, was designed to clarify the association of the GSTP1 Ile105Val polymorphism with gynecological cancer risk. Our results suggested that there was no significant association between the GSTP1 Ile105Val polymorphism and the risk of gynecological cancer in all genetic models (GG  Gynecologic malignancies are a major cause of cancer-related death in women worldwide. Cervical, endometrial, and ovarian cancer are the second, fourth, and fifth most common types of cancer, respectively, among North American women . The exaGlutathione S-transferases (GSTs) are a superfamily of phase II enzymes that are expressed in many tissues and are responsible for the metabolism of various xenobiotics and carcinogens by catalyzing the conjugation of glutathione to electrophilic compounds. They play a vital role in protecting the cells from oxidative damage and in modulating the induction of other enzymes and proteins in response to DNA damage; therefore, they are important for maintaining genomic integrity . NumerouIn the past decade, several case-control studies about the relationship between the GSTP1 Ile105Val polymorphism and gynecological cancer susceptibility have been conducted. However, the results are still inconclusive, owing to the relatively small sample sizes of the studies, resulting in limited power to estimate the overall effects. Meta-analyses are powerful tools for summarizing inconclusive results from individual studies. Therefore, we performed the present meta-analysis using data from 10 previously published case-control studies –13 to moAs shown in Figure For the overall study population, our study did not detect a significant association between the GSTP1 Ile105Val polymorphism and gynecological cancer risk in any of the genetic models .Funnel plots were created and Egger's test was performed to assess the publication bias of the included studies. The funnel plots did not show obvious asymmetry in the overall population. The results of Egger's linear regression test confirmed the funnel plot symmetry  using combinations of the following search terms: ‘GSTP1,’ ‘glutathione S-transferase P1,’ ‘polymorphism,’ ‘gynecologic malignancies,’ ‘ovarian cancer,’ ‘cervical cancer,’ and ‘endometrial cancer.’ Moreover, additional eligible articles were identified through scanning of the references cited in the retrieved articles. We assessed the titles and abstracts of the retrieved articles to exclude any irrelevant studies. Additionally, the remaining full-text articles were thoroughly scrutinized to determine whether they were suitable for the final analysis.The inclusion criteria were as follows: (1) case-control studies that were conducted to assess the association between the GSTP1 Ile105Val polymorphism and gynecological cancer risk and (2) studies with sufficient data for estimating odds ratios (ORs) and their 95% confidence intervals (CIs). Gynecologic malignancies were defined as cervical cancer, endometrial cancer, and ovarian cancer. In cases of multiple publications using the same or overlapping data, we chose the one with the largest number of subjects. The exclusion criteria were as follows: (1) reviews and comments; (2) studies with duplicate data; and (3) articles with no available information on the genotype frequency.The data were obtained according to a standardized data extraction form. In order to minimize the risk of bias, two investigators extracted the following data from each study independently: the first author's name, year of publication, country, sources of controls , ethnicity (Asian or Caucasian), cancer type , smoking status (smokers or nonsmokers), numbers of different genotypes in the case and control groups , the Hardy-Weinberg equilibrium of the controls, and the minor allele frequency in the controls.The association between the GSTP1 Ile105Val polymorphism and the risk of gynecological cancer was estimated by using ORs with their corresponding 95% CIs. Pooled ORs were obtained for comparisons of heterozygotes (GA vs. AA), homozygotes (GG vs. AA), the dominant model (AG/GG vs. AA), and the recessive model (GG vs. AA/AG). Moreover, the studies were stratified and analyzed by ethnicity, cancer type, and smoking status. The Cochrane Q statistics test and I-squared (I2) metric were used to test heterogeneity. When the I2 for heterogeneity was >50%, the fixed effects model was used. Otherwise, the random effects model was applied (DerSimonian and Laird method) . Funnel


We performed this meta-analysis to further assess the polymorphisms of platelet membrane glycoproteins with a risk of CAD.  Recently, controversial results have been reported regarding the association of the polymorphisms of platelet membrane glycoproteins . Analyses were performed using the Cochrane software package Review Manager 5.2 and Stata 12.0 software package. α VNTR polymorphism was associated with CAD in any of the genetic analysis models.  Twenty-nine full-text articles were included in the meta-analysis. Based on random-effects meta-analysis, a significant association between the HPA-2a/b polymorphism and CAD was identified . Our study showed no association between the GP VI T13254C polymorphism and CAD in either a random-effects model or a fixed-effects model. Furthermore, there was no evidence to suggest that the GP Ibα VNTR were associated with CAD. The HPA-2a/b polymorphism correlated significantly with a risk of CAD, and the HPA-2b allele and the HPA-2ab + HPA-2bb genotype may increase the risk of CAD. There was no evidence to suggest that polymorphisms of GP VI T13254C and GP Ib Coronary artery disease (CAD), one of the leading causes of mortality, is a complex disease that is multifactorial and clearly influenced by environmental factors and genetic predisposition. Platelets and thrombosis play a pivotal role in the pathogenesis of atherogenesis in CAD . Plateleα VNTR) polymorphisms with the risk of CAD. The present meta-analysis attempts to use all relevant evidence to answer the question of whether the polymorphisms of platelet membrane glycoproteins HPA-2a/b, GP VI T13254C, and GP Ibα VNTR are associated with a risk of CAD.Many studies have focused on exploring the correlation between polymorphisms of platelet membrane glycoproteins and CAD; but unfortunately there has not been some kind of unified understanding –9. So faα VNTR) with a risk of CHD by searching the Embase, Cochrane Library, PubMed, and CBM databases from their inception through December 31, 2016. The following search terms were used: “platelet glycoprotein” or “GP I b-IX Complex” or “GP I bα” or “platelet membrane glycoprotein VI” or “HPA-2” or “GP I bαVNTR” or “GP6” AND “polymorphism or gene or variant or mutation” AND “coronary heart disease” or “CAD” or “myocardial infarction” or “angina” or “ischemic heart disease” or “ischemic cardiomyopathy” or “coronary artery stenosis”. We also reviewed the reference lists of the retrieved articles and recent reviews.We initially identified published studies that investigated the association of polymorphisms of platelet membrane glycoproteins  the odds ratios (ORs)/risk ratio  and the corresponding 95% confidence intervals (CIs), or the distributions of genotype and allele frequency were reported. Two authors independently evaluated the eligibility of all studies retrieved from the databases. Disagreements were resolved by discussion or in consultation with a third author.We included any study that met all of the following criteria: (1) the study was a cross-sectional, case control, or cohort design; (2) clear diagnostic criteria for CAD were established; (3) the association between of HPA-2a/b, GP VI T13254C, and GP IbTwo investigators independently extracted data on patient and study characteristics and the quality of each study using a standardized data extraction form. Any discrepancies were resolved by consensus. The following information was extracted from each study: first author, year of publication, location, race, diagnostic criteria, gene, mean age, disease definition, gender, source of controls, number of patients, effect size , and adjusted variables.t-test. Qualitative data were compared by the chi-square test. Allele frequencies were estimated by the gene counting method, and the chi-square test was used to identify significant departures from the Hardy-Weinberg equilibrium. The strength of the association between the HPA-2a/b, GP VI T13254C, and GP Ibα VNTR polymorphisms with risk of CAD was measured by odds ratios (ORs) and 95% confidence intervals (CIs).Statistical analyses were performed using the Cochrane software package Review Manager 5.2 and the Stata 12.0 software package. Descriptive statistics were calculated for all demographic and clinical characteristics of the study subjects. Quantitative clinical data were compared between patients with CAD and control subjects using an unpaired Student's The OR values of the two-allele polymorphism were calculated from the following two models, respectively: (1) allele model ; (2) dominant model (heterozygous + homozygous mutant versus homozygous wild-type). The OR values of the variable number of tandem repeat (VNTR) polymorphism were calculated from the following three models, respectively: (1) allele model ; (2) dominant model ; (3) codominant model (single genotype versus other genotypes).I2 statistic. Where the I2 value was 50% or lower, indicating no evidence of heterogeneity, the fixed-effects model was suggested; otherwise, we used the random-effects model. In the presence of heterogeneity, we performed subgroup and sensitivity analyses to explore possible explanations for the heterogeneity and examine the influence of various exclusion criteria on the overall risk estimate. The funnel plot and Egger's test were conducted to identify a potential publication bias.Heterogeneity was quantified using the α VNTR, simultaneously [A flowchart of the study selection process is shown in aneously , 11, 16.aneously , 21, 22,aneously   and dominant model   . Cumulat = 0.02) . Howeverα VNTR polymorphism. The overall results from the different genetic analysis models identified no significant association between GP Ibα VNTR polymorphism and CAD  or the dominant model  . In this and CAD .I2 = 78%, dominant model I2 = 84%). Heterogeneity analysis revealed that the study of Chen and Zen from East Asians ethnic studies was the largest source of heterogeneity (P ≥ 0.05). This shows that the heterogeneity is derived from the statistical methods. We also found significant heterogeneity in the meta-analysis of the association between the GP VI T13254C allele model, the GP Ibα VNTR polymorphism, and CAD. Because less than 10 studies were included in the meta-analysis, a metaregression analysis was not performed to explore the source of heterogeneity.A significant heterogeneity was present in the HPA-2a/b models  and 1.30 (1.03–1.65), respectively, even after exclusion of the largest sources of heterogeneity from Chen and Zen. This fully shows that the meta-analysis results for the association between HPA-2a/b polymorphism and CAD have considerable stability. Meanwhile, this indicates that the result of Chen and Zen is the main factor leading to the increase in heterogeneity. Similarly, a sensitivity analysis did not change the results of the meta-analysis in the allele model or the dominant model for GP VI T13254C and the different genetic analysis models for GP IbP = 0.767; HPA-2a/b dominant model: P = 1.000; GP VI T13254C allele model: P = 1.000; GP VI T13254C dominant model: P = 0.707; GP Ibα VNTR: omitted)  .α VNTR were associated with CAD.A total of twenty-five studies were included in our systematic review. The results indicated that the HPA-2a/b polymorphism was associated with CAD in both the allele model and dominant model. There was no evidence to suggest that polymorphisms of GP VI T13254C and GP IbA1/PlA2 (HPA-1a/b) polymorphism of the platelet membrane glycoprotein receptor IIIa. Wu et al., Di Castelnuovo et al., and Burr et al. demonstrated meta-analysis results supporting a GP III a PlA1/PlA2 polymorphism associated with cardiovascular disease or coronary heart disease, although this correlation was rather weak. But in the years that followed, Ye et al. and Bentley et al. showed a meta-analysis whereby a platelet membrane glycoprotein polymorphism of PlA1/PlA2 (including GP Ibα Kozak-5T/C and GP Ia 807C/T) was associated with coronary heart disease, which was not statistically significant. This shows that the criteria for literature selection and meta-analysis method have a great influence on the results.At the beginning of this century, a number of researchers conducted systematic reviews on the association of platelet membrane glycoprotein polymorphisms and risk of CAD. One of the most studied was the Plα VNTR polymorphism using more complex gene analysis models).The main differences in the research methodology between this study and the meta-analyses reported above are as follows: (1) this study is limited to case control studies, and any retrospective or cohort studies were not included. (2) Studies were excluded that did not meet the Hardy-Weinberg equilibrium. (3) We used the Newcastle-Ottawa Scale (NOS) to evaluate the quality of the papers; this scale is mainly applied to a quality evaluation of the literature in case control and cohort studies. (4) A variety of genetic analysis models were used in combined-effects models (allele model, dominant model, and GP IbαVNTR, which only included 5 papers. These weaknesses can only be compensated for by carrying out more comprehensive literature collection and more rigorous clinical trials.This study still has some shortcomings. We failed to collect the unpublished literature, which is a type of study likely to avoid a publication bias. Due to the statistical analysis of a small part of the original documents using only the dominant model, no specific or calculated allele data, and the fact that we contacted relevant authors through e-mail and oftentimes did not receive a reply, a combined analysis of the allele model could not be carried out. The very strict inclusion and exclusion criteria resulted in a final statistical analysis of a smaller portion of the research, especially for GPIbα VNTR polymorphisms was related to an increased risk of CAD. However, in the future, we will need a larger sample size, a more appropriate population, and better genetic testing methods to verify this view.We confirmed that the HPA-2a/b polymorphism is significantly associated with CAD and the HPA-2b allele and HPA-2ab+bb genotype are the genetic risk factors for CAD. We found no evidence to support the theory that the presence of GP VI T13254C and GP Ib


Subsequently, we conducted subgroup analysis for 5-HT2A T102C polymorphism based on ethnicities and APOE ε4, and identified a significantly increased risk for the allelic and dominant models of 5-HT2A T102C and AD in Asian subgroup  and subgroup without APOE ε4 . Nevertheless, the pooled analyses suggested no significant association between allelic, dominant, and recessive models of 5HTTLPR (L/S) and AD (p > 0.05). In conclusion, our meta-analysis demonstrates that 5HT2A C10T, but not 5HTTLPR (L/S), might increase risk for AD.Alzheimer’s disease (AD) is a progressive neurodegenerative disease. Relationships of 5HT2A and 5HTTLPR polymorphisms and AD risk have been widely investigated previously, whereas results derived from these studies were inconclusive and controversial. The aim of this study was to investigate the association of the 5-HT2A and 5HTTLPR polymorphisms and AD using a meta-analysis of existing literatures. Studies were collected using PubMed, Web of Science, the Cochrane Library databases, Chinese National Knowledge Infrastructure (CNKI) and Embase. Pooled odds ratios (ORs) with 95% confidence intervals (CIs) were used to assess associations. As a result, a total of 7 publications for 5-HT2A T102C and 16 publications for 5HTTLPR (L/S) comprised 3255 cases and 3690 controls fulfilled the inclusion criteria. Significant association was covered between allelic and recessive models of 5-HT2A T102C and AD (allelic model:  Alzheimer’s disease (AD), characterized by progressive memory and language impairment, cognitive deficits, and other behavioral and psychological symptoms of dementia (BPSD), is a progressive neurodegenerative disorder –2. The mSerotonin  is a key neurotransmitter involved in many aspects of psychological processes including mood, aggression, impulsivity, and anxiety in human and animal –13. SeroPolymorphisms in the serotonin-related genes were demonstrated to be associated with the risk of AD in recent studies. The most commonly and widely studied polymorphisms should be the 5-HT2A (C102T) and SLC6A4 (5HTTLPR) –23. The In light of these controversial and inconclusive observations, we conducted a meta-analysis to investigate the possible role of 5-HT2A (C102T) and 5HTTLPR (L/S) polymorphisms in susceptibility of AD.As shown in Figure p = 0.08) of 5HT2A C102T   and AD are listed in Table p > 0.05) .For 5HTTLPR (L/S), a total of 16 case-control studies containing 2244 cases and 2842 controls were involved. Nevertheless, the pooled ORs for the allelic contrast , dominant model  and recessive model  of 5HT2A C102T failed to show significant associations Table . Further5) Table . In addip > 0.05) ; recessive model: Asian: NA; APOE ε4+: NA) heterogeneity and the result remained none significant, which showed that it had the highest effect on the correction of 5HT2A and AD. Furthermore, subgroup analysis stratified by ethnicities and APOE ε4 was performed and showed no obvious difference , implying that the ethnicity and APOE ε4 exerted no influence on the association between the 5HT2A C102T polymorphism and risk of AD.Considering the great heterogeneity among studies, the random-effect model was applied. Heterogeneity was found for the 5HT2A allelic and recessive models of 5HT2A C102T and AD in Asian subgroup and subgroups without APOE ε4 (Table p > 0.05). And subgroup analysis stratified by ethnicity and APOE ε4 was performed and showed no obvious difference , implying that the ethnicity and APOE ε4 exerted no influence on the association between the 5HTTLPR (L/S) polymorphism and risk of AD.Significant heterogeneities were also found in allelic, dominant, and recessive models of 5HTTLPR (L/S) Table . The hetegger=0.955) and 5HTTLPR L (pegger=0.924)  was associated with AD.The 5HT2A gene, which codes for the serotonin receptor type 2A, is located at 13q14-q21 . The C/CA functional polymorphism in the 5- regulatory promoter region, termed 5-HTTLPR, has been investigated in psychosis, mood disorder, BPSD, affective disorder, and AD –56. 5-HTNonetheless, limitations also need to be acknowledged in our meta-analysis. Firstly, we enrolled a particularly small number of studies analyzing for association between the 5HT2A C102T and AD (7 case-control studies), which may result in an insufficient power for identifying relationship of 5HT2A C102T and AD risk. Secondly, we involved only Asian and Caucasian populations in the present study. Other populations such as African were not included. However, we could not assess the association in African population for lack of studies. Therefore, future studies on various ethnicities are needed. Thirdly, further subtle adjusted analysis by other co-variants such as ages, gender, education level, and life style should be carried out to obtain a more precise evaluation. Fourthly, AD was a progressive neurodegenerative disease with age and gender bias. It is necessary to analysis the genetic association between the 5HT2A (C102T) and 5HTTLPR (L/S) in subgroups stratified by age or gender.In conclusion, our meta-analysis suggests that 5HT2A C102T may increase susceptibility to AD in Asian population and subgroup without APOE ε4 in both allelic and dominant models. And, the 5-HTTLPR (L/S) might not be the risk factor for AD. However, large-scale studies with more subjects are warranted to confirm these findings.This meta-analysis followed the Cochrane collaboration definition and PRISMA 2009 guidelines for meta-analysis and systematic review. Literatures search on PubMed, Embase, Web of Science, the Cochrane Library databases and Chinese National Knowledge Infrastructure (CNKI) was performed to investigate all relevant publications exploring the relationship between 5HT2A and 5HTTLPR polymorphisms and the risk of AD . The search terms were following: “5HT2A” or “neurotransmitter 5 hydroxytryptophan 2A Receptor” or “serotonin receptor 2A” or “serotonin 2A Receptor” or “HTR2A” and “polymorphism” or “variant” or “gene mutation” “single nucleotide polymorphism (SNP)” or “gene variation” and “Alzheimer’s disease” or “AD” and “promoter region of the serotonin transporter gene” or “5HTTLPR”. No language was limited. Meanwhile, other potentially relevant literatures were identified by cross-references within eligible studies.1) Investigating the association between 5HT2A (C102T), 5HTTLPR (L/S) polymorphisms and susceptible of AD. 2) The study was case-control and/or cohort designed. 3) Sufficient published data for calculating an odds ratio (OR) with 95% confidence interval (CI). 4) The genotype distributions in control groups were in the Hardy-Weinberg equilibrium (HWE).1) Duplicated studies, abstracts, letters or reviews. 2) Studies without controls. 3) Control group did not confirm to Hardy-Weinberg equilibrium (HWE). 4) No available genotype data.Data abstraction was performed independently by L. T. and Y. W. The following information from each study was summarized: first author, year, ethnicity, numbers of cases and controls, mean age and gender, methods of genotyping, sample source, Hardy-Weinberg equilibrium (HWE) for control groups. All included studies were evaluated using the Newcastle-Ottawa Scale (NOS) independently by L. T. and Y. W. C. Any discrepancies in the assessment were resolved by J. M. L.2-based Q test and the I2 statistic. When there was no significant heterogeneity across studies (I2 < 50%), the fixed effect model (Mantel-Haenszel method) was used for meta-analysis. Otherwise, the random effect model (the DerSimonian and Laird method) was used. Sources of heterogeneity were evaluated by stratification analysis of ethnicities and APOE ε4 allele, according to the study characteristics. Sensitivity analysis was performed to assess the stability of results. The publication bias was detected with Begg’s test and Egger’s test. p < 0.05 was considered statistically significant.The odds ratio (OR) and 95% confidence interval (95% CI) were calculated for evaluating the association between 5HT2A T102C, 5HTTLPR L/S and AD using the RevMan 5  and STATA12.0 . The pooled ORs were calculated in the allelic, dominant and recessive models. The statistical significance of the OR was determined using the Z test. Statistical heterogeneity was tested using χ


The above results suggested that there was no significant association between PMH and the risk of nephrolithiasis. The difference in menopausal status did not influence the relationship between PMH and the risk of kidney stone formation. Menopause is reported to be associated with increased urinary calcium excretion, which may enhance the risk for the development of calcium kidney stones. However, it remains controversial about whether high level of postmenopausal hormone (PMH) is a risk factor for formation of nephrolithiasis. Several observational studies have shown that PMH is protective based on 24-hour urinary parameters. Recent clinical trials provided evidence to conclude that estrogen therapy increases the risk of nephrolithiasis in healthy postmenopausal women. Our study aimed to comprehensively assess clinical evidence on the relationship between postmenopausal hormone level and risk of nephrolithiasis. To conduct systematic review, we pooled total 98 potentially related articles in Cochrane library, Medline, and Embase. Three studies with a total of 71101 study participants that included two clinical trials, 4 stratified and potentially usable results by the status of menopause and type of PMH use derived from one prospective cohort study, and one case-control studies were selected to pool relative risk using random-effect model. How the difference in menopause status, whether naturally menopausal or surgically menopausal, influenced the pooled relative risk was included in the subgroup analysis. The study population aged from 45 to 70 years old. The follow-up year and adjusted confounders differed across different studies. The pooled relative risk for the 7 stratified studies was 0.91 : ). In the menopausal status-specific analysis, the pooled relative risk for naturally menopausal women was 0.92 (95 % CI, ; I Nephrolithiasis (kidney stone) is a common disease among postmenopausal women, affecting around 5 %-7 % of the population in the United States . A sharLiterature searching was performed by using Cochrane library, Medline, and Embase. The searching terms included combinations of keywords and their synonyms, such as “Nephrolithiasis”, “postmenopausal”, “Estrogen”, “kidney stone”, “renal lithiasis”, “renal stone”, “women”, “observational”, and “Clinical Trials”. And the search strategies used for the other databases were similar, with the necessary adaptions made. Additionally, we manually searched reference lists in the selected studies to identify potentially relevant studies.We set up the inclusion and exclusion criteria to include eligible studies. We focused on postmenopausal women aged around 45 to 79 years with no previous history of kidney stones and selected cohort studies that compared the risk of kidney stones among the group with PMH treatment and the group without. The estimated values of odds ratio (OR), relative risk (RR) or hazard ratio (HR) and its 95 % confidence interval (CI) needed to be specifically reported. Demographic characteristics such as age, sex, body mass index (BMI), serum calcium, and urinary calcium were described. Confounders such as age were adjusted if necessary. The exclusion criteria included: (1) only age-adjusted or other confounder-adjusted OR, RR, or HR were reported. (2) The subjects of the study had recurrent nephrolithiasis or other significant chronic kidney diseases. (3) The data were extracted from the same study population. After carefully reviewing the selected studies, the following information on the publications was collected: abstract, full text, title, author information , and publication year. Characteristics included country of origin, sample size, age, follow-up time, type of risks and confounders-adjusted risks were extracted from the selected studies. Multivariate-adjusted outcome data  were extracted and converted logarithmically to log (RR) for each study. The I-squared statistic was used for heterogeneity test, and the log (RR) of the studies were pooled using fixed effect models when the heterogeneity is low . Otherwise, a random effect model was used. Forest plots were constructed to visually demonstrate RRs and their confidence interval. Funnel plot was constructed after Trim-and-Hill correction for evaluating funnel asymmetry. Fail-safe N was used to determine the number of NULL studies that have to be added to reduce the significance of the meta-analysis to (0.05). Publication bias was analyzed by using Egger's regression test. The sensitivity analyses were conducted by omitting one study at a time to recalculate the pooled RR. All above analyses were performed by R version 3.3.2 with functional packages including ggplot2, metafor, rmeta, and meta.A total of 98 potentially related articles were identified after employing the searching strategy in Medline (PubMed), Embase, and Cochrane library. 76 duplicated articles or non-research type's articles or non-English written articles were removed after screening their titles and abstracts. The remaining 22 articles were carefully reviewed by several independent reviewers. The selection process was demonstrated as shown in Figure 12, while mean age and BMI of surgical postmenopausal females were 58.3 and 25.8. The study by Maalouf et al.  containal. (2013) were na2 is 78.22 %, suggesting a high inter- and intra-study variation. Thus, a random effect modeling was used to perform the meta-analysis. According to the result analyzed by random effect modeling , into tFigure 2A, there iFigure 3A-B2 = 82.74 %; Surgically postmenopausal women: pooled RR, 0.90; 95 % CI, ; I2 = 78.47 %). In our subgroup analysis, we aimed at investigating whether different menopause types influenced the association between PMH and the risk of nephrolithiasis. According to the results  did not contribute to identifying significantly increased or decreased risk of kidney stone.To investigate whether the selected studies showing a more significant intervention effect than studies with null results. We plotted Funnel plot with the Trim-and-Fill adjustment and further conducted Fail-safe N method. Null study was needed to nullify the effect. We observed in the plot that the studies with higher precision tends to report higher relative ratio. Subsequently, we performed Egger's regression test to analyze funnel plot asymmetry Figure 5. Based oThe incidence of kidney stones is increasing worldwide. Kidney stones were reported to affect approximately 1 in 11 people in the United States. The prevalence of kidney stones was 8.8 % . KidneyMenopause status is associated with increased urinary calcium excretion, which may increase the risk for the formation of calcium-containing kidney stones . HormonOur study has several limitations. Due to limited publications reporting the association between PMH and formation of kidney stone, only 3 studies were included in this meta-analysis. In the study of Mattix Kramer et al. (2003) lackingNone.The authors report no conflicts of interest. The authors alone are responsible for the content and writing of the paper.


We conducted a systematic review and meta‐analysis of interferon regulatory factor 6 and 8q24 polymorphisms with nonsyndromic cleft lip with/without cleft palate (NSCL/P).Data extraction was independently performed by two reviewers. Genotypic effects of four polymorphisms from 31 studies were pooled separately by ethnicity using a mixed‐effect logit model with accounting for heterogeneity.For rs2235371, AA and GA carried, respectively, 51%  and 42%  lower risks of NSCL/P than GG genotypes in Asians, but these genotypes were not significant in Caucasians. For rs2013162, only AA was significant, that is, carried 0.65  times lower odds than CC in Caucasians but not for Asians. For rs642961, AA and GA genotypes, respectively, carried 2.47  and 1.40  times higher odds in Asian, and 2.03  and 1.58  times higher odds in Caucasians compare with GG genotypes. For rs987525, AA and CA genotypes carried 2.27  and 1.34  times higher odds in Asian, and 5.25  and 2.13  times higher odds in Caucasians, and 1.42  and 1.28  times higher odds in mixed ethnicities compared with CC genotypes. These variant effects remained significant based on applying Bonferroni corrected‐thresholds, except in the mixed ethnicity.We show robust variant effects in NSCL/P. Considering them with other genes and risk factors might be useful to improve prediction of NSCL/P occurrence. Birth Defects Research (Part A) 106:773–788, 2016. © 2016 The Authors Birth Defects Research Part A: Clinical and Molecular Teratology Published by Wiley Periodicals, Inc. Nonsyndromic cleft lip with or without cleft palate (NSCL/P) is one of the commonest birth malformations and varies by geographic origin and ethnicity Bender, . The incPatients with NSCL/P suffer from functional and cosmetic problems, and NSCL/P can affect speech and communication, which results in delayed development, low self‐confidence, and poor quality of life for both patient and family members. Although reconstruction with maxillo‐facial surgery is possible, severe cases sometimes require multiple corrective surgeries , dental work, and speech rehabilitation. Thus, this condition places a significant burden on family and medical services Strauss, .IRF6), located on chromosome 1q32.2 (OMIM # 607199), is associated with the autosomal‐dominant Van der Woude syndrome, the most common Mendelian syndrome that has the cardinal signs of cleft lip with or without cleft palate (CL/P) and/or cleft palate only (CPO) with dental anomalies and pitted lips  was the most significantly associated with NSCL/P and was the major susceptibility locus in the general population  and the 8q24 region (at rs987525) on NSCL/P, and their effects were later pooled by a meta‐analysis  of allele/genotype effects on outcome along with its 95% confidence interval (CI). The studies with insufficient information were excluded if authors did not provide data after two contacts. Where there were multiple publications with the same or overlapping subjects  the most complete and/or recent results were used. The reference lists of the selected articles were also reviewed to identify additional relevant publications.Two reviewers (K.W. and S.R.) independently selected studies by looking through all titles and abstracts of the identified studies. Any human population‐based  association study published in English was included if it met the following criteria: the outcome of interest was NSCL/P, the studied polymorphisms were in IRF6 and 8q24 were extracted independently by two reviewers (K.W. and S.R.) using a standardized data extraction form. Data on co‐variables, for example, mean age, percentage of males, ethnicity, and family history were also extracted. The main data (generic data and risk of bias assessment) were computerized separately by reviewer. Then, the two data were compared and validated to identify disagreement between. Meetings between the two reviewers (K.W. and S.R.) and the third party (A.T.) were then setup to discuss and solve disagreement.Summary data for The quality of studies was independently assessed by 2 reviewers (K.W. and S.R.) using a risk of bias assessment for genetic association studies, described in detail previously , and one located within 8q24 (rs987525 C>A).Our review focused on NSCL/P as the outcomes of interest, which diagnosed according to original studied. The studied genes were three polymorphisms within the Data analysis was performed following a standard method of meta‐analysis for genetic association studies ; otherwise a fixed‐effect model with inverse‐variance method was applied.Variant effects on NSCL/P were assessed using a per‐allele and per‐genotype approach. For the per‐allele approach, an OR of a versus A, if a and A were, respectively, minor and major alleles, along with its variance were estimated. Heterogeneity was then assessed using Cochrane's Q test and 1 (aa vs. AA) and OR2 (Aa vs. AA) were estimated for each polymorphism across included studies. Heterogeneity was then assessed using the same methods described above. If any of the two ORs was heterogeneous, data were then pooled using a mixed‐effect hierarchical logit model , which was then estimated using the model‐free Bayesian approach , which resulted in a Bonferroni corrected threshold of 1.85 × 10‐3.Analyses were performed using STATA version 14  for 8q24 had sufficient numbers of studies; therefore, further pooling was focused on only these polymorphisms from 31 studies.A total of 307 and 1155 studies were, respectively, located from Medline by means of PubMed and Scopus Fig.  had insuIRF6, and rs987525 for 8q24, with the outcome NSCL/P, respectively.The characteristics of these 31 studies  had low risk of selective reporting of outcomes. Twenty‐seven (87%) and 24 (77%) studies described control ascertainment and quality control for genotyping, respectively.p = 1.27 × 10‐9), which was still significant based on Bonferroni corrected‐threshold. The MAF A in Caucasians was very rare and much lower than in Asians, that is, 0.03  and 0.02  in control and NSCL/P groups, respectively. This yielded a pooled OR of 0.69 , but this was not statistically significant.Fifteen studies  assessed effect of rs2235371 polymorphism on NSCL/P. Allele frequencies in the NSCL/P and control groups are described, and genotype frequencies in the control groups complied with HWE for all studies (Supplementary Table 4). The MAF was pooled by ethnicity. This suggested the pooled prevalence of MAF A in Asians was 0.34  in the control group and 0.26  in NSCL/P groups, yielding a pooled OR of 0.66  and control (n = 4756) groups along with ORs for all studies are described in Table 2I for OR1 (AA vs. GG) and OR2 (GA vs. GG) were, respectively, 60.0%  and 0%  for pooling 9 Asian studies (n = 1677 vs. 1717). The corresponding pooled ORs were 0.49  and 0.58 , see Figure The genotype frequencies between NSCL/P (n = 1704 and 2669 for NSCL/P and control), heterogeneity was not present in OR1 but was present for OR2 with an 2I of 0%  and 51.9% , respectively. The pooled OR1 and OR2 were 0.75  and 0.82 , see Figure Among 5 Caucasian studies  and OR2  in Asian, and OR1  and OR2  in Caucasian.Publication bias was assessed for both pooled ORs in Asian and Caucasian by funnel plot and Egger's test. There was no evidence of asymmetry of the funnel, see Supplementary Figure 1A–D. These results agreed with the Egger's tests, which indicated no evidence of asymmetry of funnels for both ORp = 4.16 × 10‐4) and 0.99  in Caucasians and Asians (Supplementary Table 5).Among the nine studies that assessed the association between rs2013162 in IRF6 and NSCL/P, four studies, three studies, and two studies were conducted in Caucasians, Asians, and in mixed ethnicity, respectively. Allele effects were pooled in Caucasian and Asian where numbers were sufficient. All studies complied with HWE with the pooled MAF A of 0.35  and 0.42 , respectively, in Caucasian, Asian of control populations. The pooled ORs of A versus C alleles were, respectively, 0.84 , ORs  showed low heterogeneity for OR1  and OR2 , see Figure 1 and OR2 were, respectively, 0.65  and 0.89 , suggesting that the odds of NSCL/P was significantly different in AA genotype even based on Bonferroni corrected threshold, but was not significantly different in CA genotype when compared with CC genotype. The estimated lambda was 0.32 , which possibly indicated between a recessive to additive effect of A allele.The genotype frequency between NSCL/P groups of 9 studies are described in Table n = 501 for NSCL/Ps and 633 for controls), heterogeneity was present in both OR1 and OR2 with an 2I of 88.4%  and 82.6% , respectively. The pooled OR1 and OR2 were 1.40  and 1.20 , see Figure Among three Asian studies (p = 0.731) for OR1 and ‐0.25  for OR2 in Caucasian, and ‐8.58  for OR1 and ‐7.41  for OR2 in Asian.Publication bias was assessed for both pooled ORs in Caucasian and Asian by funnel plot and Egger's test. Funnel plots suggested little asymmetry , but the Egger's test did not confirm this. The coefficients of asymmetry was ‐2.07  and 1.47  in Caucasians and Asians, respectively.Eleven studies assessed effect of rs642961 polymorphism on NSCL/P. Among them, five studies, four studies, and two studies were conducted in Caucasians, Asians, and mixed ethnicity, respectively. One study , the OR1 (AA vs. GG) and OR2 (GA vs. GG) were estimated with low heterogeneity for OR1  and moderate for OR2 . The mixed‐effect logit model yielded pooled OR1 and OR2 of 2.03  and 1.58 , which suggested that persons with AA and GA genotypes were 2.03 and 1.58 times significantly higher risk of NSCL/P than persons with GG genotype , the OR1 (AA vs. GG) and OR2 (GA vs. GG) were estimated with moderate heterogeneity for OR1  and OR2 . The mixed‐effect logit model yielded pooled OR1 and OR2 of 2.47  and 1.40 , which suggested that persons with AA and GA genotypes were 2.47 and 1.40 times significantly higher risk of NSCL/P than persons with GG genotype  suggesting a potential mode of variant effect could be from a recessive to additive mode.Among four Asians studies  and funnel plots .No evidence of reporting bias was seen with either ORp = 1.21 × 10‐4), and 2.20 , and 1.21 , respectively.Allele frequencies of 12 studies assessed effect of rs987525 polymorphism on NSCL/P and all observed HWE (Supplementary Table 7). The MAF A in controls were 0.15  in three Asian studies, and 0.20  in four Caucasian studies, and 0.27  in five mixed ethnicity. The estimated ORs for allele A versus C in these corresponding ethnicities were 1.48 , the pooled OR1 (AA vs. CC) and OR2 (CA vs. CC) were, respectively, 2.27  and 1.34 , with no heterogeneity for OR1  and OR2  (Table Among three Asian studies (n = 955 in NSCL/P and 8735 in controls), the pooled OR1 (AA vs. CC) and OR2 (CA vs. CC) were, respectively, 5.25  and 2.13 , with no heterogeneity for OR1  and moderate heterogeneity for OR2  (Table Among four Caucasian studies (n = 1370 in NSCL/P and 1719 in controls), the corresponding pooled ORs were 1.42  and 1.28  with moderate heterogeneity for OR1  and no evidence of heterogeneity for OR2   for IRF6 and for rs987525 in 8q24. The MAF for these corresponding polymorphisms were similar across ethnicities except for rs2235371 where the MAF was common in Asians but rare in Caucasians, whereas MAF for rs987525 was common in Caucasians but less common in Asians. Because the MAF of rs2235371 was different between ethnicities, the genotype effect for this SNP was different. All polymorphisms were significantly associated with NSCL/P in some ethnicities. A significant protective effect of rs2235371 in Asians and effect of rs2013162 in Caucasians were suggested, that is, carrying one copy of the A allele for both polymorphisms lowered the odds of NSCL/P compared with G and C alleles, respectively. In addition, our findings indicated risk effects of minor alleles for rs642961 and rs987525.We conducted a systematic review and meta‐analysis to assess genetic effects of polymorphisms in IRF6, and rs987525 for 8q24, respectively. We confirmed the previous findings by expanding to a larger number of included genetic association studies  and applying Bonferroni correction for multiple tests. Although each polymorphism was considered separately in meta‐analysis for genetic association studies, this type of study had been used to replicate and confirm GWAS. Not only allele effects were estimated, but also pooled MAF and genotype effects by ethnicity additionally provided from previous evidences.Our findings are consistent with previous human GWAS studies  and 8q24 (rs987525) SNPs contributed to NSCL/P, in which the rs987525 had highest contribution. This might be comparable to the effect of folate with an estimated PAR approximately 18%  but not for rs642961 . To determine this required raw data for performing a haplotype analysis. Some variant effects were heterogeneous  across studies; we could explore for a cause of heterogeneity by fitting percent male in a meta‐regression, but this was not detected (data were not shown). There might be other characteristics that cause heterogeneity, but those data were not available. Finally, there might be gene to gene interactions, but summary data do not allow us to explore this.Our study has some strengths. We attempted to pool variant effects on NSCL/P separately by ethnicity if data were available. Prevalence of MAFs for each study polymorphism was pooled across studies and ethnicity. Mode of variant effect was estimated using a mixed‐effect logistic regression  to capture lambda (Minelli et al., IRF6 and rs987525 in 8q24 were significantly associated with NSCL/P. However, after applying Bonferroni corrected‐thresholds, variant effects remained significant, except in mixed ethnicity. We show robust variant effects in NSCL/P. Including them with other risk factors and other genes in a risk prediction model of NSCL/P might be useful to improve prediction of NSCL/P occurrence.All polymorphisms, rs2235371, rs2013162, and rs642961 at Additional Supporting information may be found in the online version of this article.Supplement Figure 1. Funnel plots for rs2235371 of IRF6.Click here for additional data file.Supplement Figure 2. Funnel plots for rs2013162 of IRF6.Click here for additional data file.Supplement Figure 3. Funnel plots for rs642961 of IRF6.Click here for additional data file.Supplement Figure 4. Funnel plots for rs987525 of 8q24.Click here for additional data file.Supplement Tables.Click here for additional data file.


Meta-analysis included 25 studies with 6334 MI patients and 6711 controls. Bioinformatics tools were applied to evaluate the impact of M235T polymorphism on angiotensinogen function and structure.P = 0.029) and T allele  and MI risk. Meta-analysis also revealed a significant association between M235T polymorphism and MI risk in allelic  and recessive  models within Asian population. In silico-analysis revealed that M235T fundamentally changed the function of angiotensinogen .Genetic association study revealed a significant association between TT genotype (odds ratio [OR] 2.08, 95% confidence interval [CI] 1.08-4.00, Our study suggests that M235T polymorphism might be a helpful biomarker for screening of susceptible individuals for MI in Asian population. AGT gene modify the plasma concentration of AGT and may be implicated in the pathogenesis of hypertension, coronary heart disease, and myocardial infarction  on exon 2 of the AGT gene. This transition results in substitution of methionine to threonine at codon 268, which is traditionally known as M235T. In recent years, several researchers have investigated the association between AGT-M235T polymorphism and MI, with inconsistent results  is a major cause of mortality, especially in industrial countries . Its prefarction ,12. Ther results -17. The This case-control study included 340 participants: 155 patients with acute MI and 185 healthy controls. We screened about 250 patients with acute MI admitted to the Coronary Care Unit of the Shahid Beheshti Hospital  between 2013 and 2015. The mean age ± standard deviation of patients was 62.37 ± 3.21 years . Patients with the history of coronary artery (n = 23), vascular (n = 12), renal (n = 7), liver (n = 3), thyroid (n = 4), diabetes (n = 34), and any other familial and genetic diseases were excluded from the study. MI was confirmed by patient history, ECG changes, increased creatine kinase (CK)-MB activity, and high levels of serum troponin (T or I). The controls were selected from people who referred to the same hospital for a routine check-up. The control group included 185 healthy participants without symptoms of coronary diseases and any other familial and genetic diseases. Control participants were free from MI as shown by medical history, electrocardiography, and clinical examination. The mean age ± standard deviation of controls was 61.68 ± 4.28 years . 2 mL ofAGT-M235T) by polymerase chain reaction restriction fragment length polymorphism (PCR-RFLP) method. For this purpose, total genomic DNA was isolated from blood samples by DNG-plus Kit . The primer sets used for amplification of AGT fragment were 5′-CCGTTTGTGCAGGGCCTGGCTCTCT-3′ and 5′-CAGGGTGCTGTCCACACTGGACCCC-3′. PCR was carried out in a total volume of 25 µL, including 2.5 µL 10X PCR buffer, 0.5 U of Taq DNA polymerase, 1.5 µM MgCl2, 0.5 µL dNTPs, 0.35 µM of each primer, and 50 ng of template DNA . PCR conditions were 94°C for 10 min, followed by 35 repetitive cycles of 94°C for 30 s, 59°C for 45 s, and 72°C for 60 s, with a final extension at 72°C for 10 min in a peqSTAR thermal cycler . PCR products were then digested by PsyI restriction enzyme (Fermentas) at 37°C for 16 h, electrophoresed on 8% polyacrylamide gel, and visualized by silver nitrate (AgNO3) staining as described by Green and Sambrook  investigation of the AGT-M235T polymorphism and MI (ii) in a case-control study; (iii) on human beings; (iv) with accessible data to calculate the odds ratios (OR) and 95% confidence intervals (CI) als (CI) . The assAGT gene was obtained from Ensembl (Accession No. ENSG00000135744) and analyzed using GeneRunner software . The coding sequence of AGT gene was determined and translated to amino acid sequence by Expasy web server (web.expasy.org/translate). The physicochemical properties of 235M normal and 235T mutant forms were obtained from ProtParam server (http://web.expasy.org/cgi-bin/protparam/protparam). The impact of M235T substitution on the protein function was evaluated by SNAP (https://rostlab.org/services/snap/) (www.predictprotein.org/getqueries) (http://genetics.bwh.harvard.edu/pph2/) (http://fasta.bioch.virginia.edu/fasta_www2/fasta_www.cgi?rm=misc1&pgm=cho). The three-dimensional structure of the AGT protein was obtained from SNPeffect 4.0 server (http://snpeffect.switchlab.org/) . The Kyte-Doolittle hydropathy pattern (The entire genomic sequence of human s/snap/) , Predictqueries) , and Polu/pph2/) . The effab.org/)  and visu2 test. The association between the MI risk and AGT-M235T polymorphism was estimated using odds ratios (ORs) and their 95% confidence interval (CIs), which were calculated by binary logistic regression. Odds ratios were also adjusted for age, sex, body mass index, smoking status, and biochemical features. A two-tailed P-value lower than 0.05 was considered statistically significant. Statistical analyses were performed using SPSS software version 19.0 .Hardy-Weinberg equilibrium (HWE) was calculated using χPheterogeneity>0.1), the random effect model was used. Otherwise, the fixed effect model was used. For sensitivity analysis, each study at a time was excluded to detect the magnitude of the effect on the overall summary estimate. Begg’s funnel plot and Egger’s test were employed to evaluate the publication bias (http://www.cebm.brown.edu/openmeta/) and Comprehensive Meta Analysis  software were used for all calculations in the meta-analysis.In the meta-analysis, we first estimated the pooled OR and 95% CI for the following five genetic models: T vs M (Allelic), TT vs MM (co-dominant), TM vs MM (co-dominant), MT+TT vs MM (dominant), and TT vs MT+MM (recessive). Values of each study were combined by random effects model  or Manteion bias ,27. The AGT-M235T genotypes was in Hardy-Weinberg equilibrium in the MI  and control  group. The genotypes and allele frequencies for the M235T in the MI and control group are shown in P = 0.029). Furthermore, T allele carriers (MT+TT) were at a high risk for MI development . Also, we observed a significant association between T allele and MI risk .The distribution of 215 articles were identified from electronic databases and the citations in potentially relevant articles. After reading abstracts and initial assessment of the articles, 42 and 141 articles were found to be duplicates and irrelevant studies, respectively. From 32 potentially relevant reports, 24 studies from 17 countries were included in the meta-analysis -17,28-46P = 0.033) and recessive  models  and recessive  genetic models AGT-M235T and MI risk in any of the 5 genetic models in Caucasian populations  models . In Asiac models . Even, aulations .Pheterogeneity<0.001, I2 = 70%), TT vs MM , MT+TT vs MM , and TT vs MM+MT  genetic models (PEgger = 0.002), TT vs MM (PEgger = 0.058), and TT vs MM+MT (PEgger = 0.015) genetic models. Meanwhile, we did not observe any publication bias in Asian population (PEgger>0.05). Also, the shape of the funnel plot showed no obvious evidence of asymmetry for Asian population (The heterogeneity test showed a true heterogeneity between studies in the T vs M (c models . Also, apulation . SensitiThe data from Protparam server revealed that the M235T substitution  changed AGT-M235T and MI. However, other studies did not observe this association (AGT-M235T and the MI risk in 2 genetic models, but in 5 genetic models in Caucasian populations there was no significant association.This study found TT genotype and T allele to be genetic risk factors for MI. To confirm this result, we also conducted a meta-analysis, which can more accurately determine the impact of a genetic polymorphism on the risk of diseases development and progression . The metociation -17. ThisAGT-235T allele and MI risk, whereas Ko et al (Different Asian populations also exhibited a great heterogeneity. For example, Chen et al , Zhu et Ko et al , FrossarKo et al , and RanKo et al  did not AGT-M235T and the MI risk, even in the subgroup analysis comprising different ethnic and control sources. However, they (There were 3 meta-analyses that reported erroneous data from the included studies -52. Sui er, they  includeder, they ,39,40,43er, they  and Wanger, they , howeverer, they  and the er, they . Wang aner, they ,39 as CaAGT gene variants were shown to be associated with hypertension. Also, hypertensive patients with different AGT variants showed significantly different plasma concentrations of AGT (AGT variant could be a possible candidate for pharmacogenomic renin-angiotensin system blockage intervention (s of AGT . Thus, Arvention .in silico tools such as ProtParam, SNAP, PredictProtein, and PolyPhen-2 servers. ProtParam calculates numerous physico-chemical properties inferred from a protein sequence without giving any additional information. PolyPhen2 scoring makes it possible to determine the effects of SNPs on protein phenotypes. It requires input data such as location and variety of amino acid substitution (in silico analyses revealed this substitution as a mutation affecting physicochemical properties, secondary structure, and AGT function.Amino acid substitutions are a common cause of the development of human inherited diseases . Therefotitution . SNAP, atitution . AlthougAGT-M235T polymorphism on MI. In the meta-analysis, there was a lack of data from African populations. Second, a lack of original data from the studies included in the meta-analysis restricted further evaluations of the potential interactions that may modulate MI risk, such as gene-environment and gene-gene.There are some limitations of our study. In the case-control study, we set the lower age limit for study entry, although cases with earlier disease onset should also be considered. In addition, we did not evaluate factors such as epigenetics, other genes, other mutations, and environmental factors, which may modulate the effects of AGT might be a genetic risk factor for MI, especially in Asian population. Also in silico analysis revealed that this substitution may affect the AGT structure and function. Future studies in larger populations should assess epigenetics, other genes, other mutations, and environmental factors.In conclusion, data from the genetic association study revealed that M235T substitution in