Studies had attempted to clarify the relation between COX2 -765G/C gene polymorphisms and periodontitis risk, but there has been no definite consensus to date. A meta-analysis was performed to further explore the relationship of COX2 -765G/C polymorphism on periodontitis risk among Chinese population.The databases of PubMed, Springer Link, Ovid, Chinese Wanfang Databases, Chinese National Knowledge Infrastructure (CNKI) and Chinese Biology Medicine were searched up to January 2017. The overall result and subgroup analysis results were combined using fixed-effect or random-effect based on the heterogeneity.Finally, 7 case–control publications including 1399 periodontitis cases and 1663 controls were identified according to the inclusion criteria. In the total analyses, COX2 -765G/C polymorphism had nonsignificant association on periodontitis risk in all models. The subgroup analyses suggested a significantly increased risk of periodontitis in studies with population-based controls and a significantly decreased risk in studies with hospital-based controls.This meta-analysis indicated that COX2 -765G/C polymorphism had significantly affect on periodontitis risk among Chinese individuals, which should be confirmed by other ethnic groups. Periodontitis, divided into chronic periodontitis (CP) and aggressive periodontitis (AP), is a multifactorial disease, involving both genetic and environmental risk factors. Previous report had confirmed that periodontal disease is at least partially determined by genetic . The facThe databases of PubMed, Springer Link, Ovid, Chinese Wanfang Data databases, Chinese National Knowledge Infrastructure (CNKI) and Chinese Biology Medicine were searched for studies examining the relation between COX2 -765G/C polymorphism and periodontitis risk up to January 2017. The search keywords were namely:  and (cyclooxygenase-2 or COX-2) and (China or Chinese or Taiwan). No restriction was imposed on search language. In addition, we also reviewed the references cited in the searched articles to look for other related studies.The following criteria were used: (1) they were case-control or cohort studies describing the association of the COX2 -765G/C polymorphism and periodontitis, (2) they provided the genetypes in cases and controls, (3) the participants were of the Chinese population. Exclusion criteria: (1) repeated literature, (2) incomplete data, (3) case-only articles, (4) review articles and abstracts, (5) participants with systematic diseases.The Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement was used in our report. Titles and abstracts of all identified studies were screened firstly. We then reviewed full articles when it is ambiguous only read to title and abstract. Data extracted from identified studies included first author’s name, publication years, type of periodontitis, the source of controls, geographical area, cases and participants, and the people with COX2 -765G/C genotypes. Hardy-Weinberg equilibrium (HWE) in controls was calculated using corresponding genotype distribution.2-test was performed to determine the Hardy-Weinberg equilibrium (HWE) of genotypes. We also calculated the heterogeneity of rare allele frequencies by χ2-test when it is available in control groups of every individual study included. The association of COX2 -765G/C polymorphism and periodontitis risk was assessed using odds ratio (ORs) and their 95% confidence intervals (CIs). The Mantel-Haenszel’s fixed-effect model was used when the between-study heterogeneity was below 50%. Otherwise, the DerSimonian and Laird’s random-effect model was used to combine the overall ORs and 95%CIs. Significance of the pooled ORs was calculated using the Z-test. Sensitivity analysis was performed to note whether a single study could influence the overall result while removing a single study at a time. Furthermore, subgroup analyses by geographical area, source of controls, and type of periodontitis were also performed. All statistical tests were calculated by the Stata 12.0 . A p-value less than 0.05 was considered to be significant.The following models were used in our report: (1) allelic contrast, (2) contrast of homozygotes, (3) recessive, and (4) dominant models. The χFigure Table We also performed a leave-one-out analysis to explore whether the sensitivity of our meta-analysis was existence. The results were essentially unchanged while removing a single individual study at a time, suggesting that the result in our study was comparatively credible and stable Fig. .Fig. 4A Periodontitis is a multifactorial disease with the risk factors including oral microorganisms  and tobaSeven case-control studies were included in this study comprising 1399 periodontitis cases and 1663 controls. Pooled result indicated that a significantly increased risk of the association between the COX2 -765G/C variants and periodontitis in studies with population-based controls and a significantly decreased risk in studies with hospital-based controls was found. Till now, there are only two meta-analyses published on the COX2 -765G/C and periodontitis , 23, oneSeveral strengths were showed in our study. First, we strictly followed the PRISMA methodology to develop inclusion criteria and exclusion criteria to reduce the possible selection bias. Second, little publication bias and other bias would be appearance while we only included Chinese populations. Third, we investigated the influence of a geographic area and the source of control on the risk of periodontitis and COX2 -765G/C. Fourth, sensitivity analysis indicated no single study had effect on the whole result, suggesting our results are reliability and stability. Several limitations should also be considered. First, this ethnic-specific meta-analysis only included data from a single ethnic group; therefore, the result obtained in our study is only applicable to Chinese populations. Second, since this meta-analysis was based primarily on unadjusted ORs and CIs, it is not possible to control confounding factors. Third, we have devised a comprehensive search strategy, but including the number of research is still relatively small. Finally, the publication bias was not evaluated due to the limitations of funnel plot.This meta-analysis indicates a significant risk between COX2 -765G/C polymorphism and periodontitis in the Chinese population. Ethnicity and control sources seem to have a confounding effect on these studies. Based on the limited number of included studies, further studies with large cases and participants are required to confirm this result.


N-deacetylase/N-sulfotransferase (heparan glucosaminyl) 3, (NDST3). This study aimed to replicate the results in a Han Chinese population and then potentially extend these findings. We performed a two-stage study to investigate the association of NDST3 with the schizophrenia and BD risk in the Han Chinese. In stage 1, a total of 632 patients with schizophrenia, 654 patients with BD and 684 healthy controls were recruited from the Shanghai region. In stage 2, 522 schizophrenia patients and 547 normal subjects were enrolled from the Hangzhou region. Then, we conducted a meta-analysis based on the present literature. In stage 1, the single nucleotide polymorphism (SNP) rs11098403 showed a significant association with schizophrenia (corrected P=0.005). The frequency of the rs11098403 G allele was significantly lower among schizophrenia patients than among the controls (odds ratio (OR)=0.68, 95% confidence interval (CI): 0.55−0.84, corrected P=0.002). No significant difference was observed in individual SNP marker genotypes or allele distributions between the BD and control groups. In stage 2, the association of rs11098403 with schizophrenia could be validated . After pooling all data from 1861 patients with schizophrenia and 2081 controls, we observed a significant association of the rs11098403 G allele with schizophrenia , with an OR=0.70 (95% CI: 0.61−0.79). Then, we performed an expression quantitative trait loci analysis to investigate the functional effect of rs11098403 on NDST3 expression in the brain. We observed a significant association of rs11098403 with NDST3 expression in the hippocampus (P=0.027), although the significance did not survive after multiple testing correction. Our findings provided preliminary evidence that rs11098403 might modify the genetic risk of schizophrenia in the Han Chinese. Further investigations are warranted to identify the precise mechanism regulating brain NDST3 expression in the Han Chinese. These results would help to explain the pathophysiological mechanism of schizophrenia.A novel susceptibility locus (rs11098403) for schizophrenia and bipolar disorder (BD) was identified in an Ashkenazi Jewish population by a recent large-scale genome-wide association study. The rs11098403 is located in the vicinity of the gene encoding  The rs11098403 SNP is located within an intergenic region in the vicinity of the gene encoding N-deacetylase/N-sulfotransferase (heparan glucosaminyl) 3, (NDST3); the 5′ end of this gene is ~308 kb from this locus.7NDST3 is expressed abundantly in the hippocampus and cerebellum,8 in which structural and functional abnormalities are consistently observed in patients with schizophrenia10 and BD.11 Therefore, this GWAS provided suggestive evidence that the NDST3 variation predisposed patients to schizophrenia and BD.The genome-wide association study (GWAS) is currently the most powerful, systematic and unbiased genetic approach to identify susceptibility variants in complex disorders such as schizophrenia and BD.12 In a recent replication study, rs11098403 demonstrated a nominally significant association with schizophrenia in a Han Chinese population.13 However, the small sample size limited the ability to draw any definitive conclusion due to the low statistical power of detection coinciding with increased rates of both false positives and false negatives.14 In this study, we first aimed to replicate the association of rs1109803 with schizophrenia and BD in Han Chinese samples from the Shanghai region. Our results replicated the positive association between rs1109803 and schizophrenia. Subsequently, we validated this positive result in an independent Chinese Han population from the Hangzhou region. To overcome the limitation of the small sample size, we performed a meta-analysis to clarify the association between rs1109803 and schizophrenia in the Han Chinese. Second, although the GWAS data indicated that the NDST3 variation possibly predisposed patients to schizophrenia and BD, to the best of our knowledge no genetic study has identified an association of NDST3 with schizophrenia and BD. Therefore, we performed an association study to address this issue. Third, a prior literature search provided evidence that the intronic SNP rs11098403 had a regulatory effect on the expression of NDST3 in postmortem cerebellar tissue.7 We validated the role of rs11098403 in NDST3 expression in the brain using expression quantitative trait loci (eQTL) analysis via an available database.Poor replication is an important issue in genetic association studies. Therefore, these studies require cautious replication validation.All subjects were of Han Chinese descent and provided written informed consent before the performance of any procedures related to this study. All procedures of this study were reviewed and approved by the Institutional Review Boards of the Shanghai Mental Health Center and local institutions. This study was performed in strict accordance with the Declaration of Helsinki and other relevant national and international regulations.We performed this study using a multi-stage association design. For the first stage, 632 patients with schizophrenia and 654 patients with BD were collected from the Shanghai Mental Health Center, Shanghai Jiao Tong University School of Medicine. Six hundred and eighty-four healthy controls were recruited from the hospital staff and students of the School of Medicine in Shanghai and then interviewed by a specialized psychiatrist using the Structured Clinical Interview for DSM-IV-TR Axis I Disorders-Patient Edition. Stage 2 included 522 schizophrenia patients and 547 normal subjects from the Hangzhou region. The patients were enrolled from the Tongde Hospital of Zhejiang Province and the Hangzhou Seventh People's Hospital. The controls were randomly selected from the general populations in Hangzhou. The demographics of the analyzed populations were presented in 18 Diagnosis and review of psychiatric case records were independently checked and verified by two experienced psychiatrists to ensure consistency. None of the control subjects had any prior or current psychiatric disorders and/or chronic physical diseases.The patients with schizophrenia or BD were recruited on the basis of previously established criteria.NDST3 may confer susceptibility to schizophrenia. In total, five selected tagging SNPs of NDST3 were genotyped to test for a possible association. To set inclusion criteria for tagging SNPs, we retrieved CHB data from the HapMap database (http://www.hapmap.org) and defined linkage disequilibrium (LD) blocks using Haploview 4.2 . Haplotype-tagging SNPs (htSNPs) were selected with an r2 cutoff >0.8 and minor allele frequency >0.2, resulting in the capture of five htSNPs . The detailed information of the selected SNPs was presented in 15 PCR amplification was performed in a volume of 25 μl containing the primer pair for each SNP. PCR primers were also used for sequencing. Sequencing results were analyzed with DNAStar , and the original sequencing chromatograms of each sample were manually checked.The principal hypothesis underlying this study is that common SNPs in 20 Ten percent of the samples were randomly selected and independently sequenced to evaluate the genotype accuracy. Here our re-sampling results were 99.91% consistent with the genotype data from the original analysis.Quality control methods were based on our previous work.14 The literature included in the analysis was selected using PubMed (http://www.ncbi.nlm.nih.gov/pubmed/) and SCOPUS (http://www.scopus.com) searching the keywords ‘NDST3 or rs11098403', ‘Han Chinese', and ‘schizophrenia, SZ or SCZ' in varying combinations. Bibliographies or citations from the retrieved articles were also checked. Reports from these searches included research published up to June, 2015. Eligible studies were included in our meta-analysis if they met all of the following criteria: (1) association studies written in English or Chinese; (2) described the genotyping method and used commonly acceptable diagnosis criteria, such as the DSM; (3) contained independent data; and (4) presented sufficient data to calculate the odds ratio (OR) with a confidence interval (CI) and P-value. The major exclusion criteria included deviation from the Hardy−Weinberg equilibrium, overlap with previous studies, or insufficient information for extraction. The authors were contacted in cases where there were queries regarding their studies. Two independent authors extracted the following data from each eligible study: last name of the first author, year of publication, and allele frequencies of cases and controls.The meta-analysis was performed in agreement with previously described methods.17 Hence, we used the brain eQTL database (http://caprica.genetics.kcl.ac.uk/BRAINEAC/), which is a large exon-specific eQTL data set covering ten human brain regions, for this analysis. The BRAINEAC database is based on 134 healthy Caucasian brain samples. More detailed information can be found in the original study.21Considering that mental illness reasonably originates from abnormal brain functions, brain samples are presumably appropriate for eQTL analysis of risk SNP(s).http://analysis.bio-x.cn).22 Pairwise LD of all pairs of htSNPs was assessed using HaploView 4.2 (Broad Institute),23 and the extent of LD was measured by the standardized D′ and r2. Power calculations were performed with Quanto 1.2.3 (http://hydra.usc.edu/GxE). To rule out population stratification between the two disorders and control populations, additional genotyping data for the six selected SNPs in this study were analyzed with the STRUCTURE software . To adjust for multiple testing, the level of significance was corrected via Bonferroni correction. For the meta-analysis, data were summarized using two-by-two tables. The Cochran χ2-based Q statistical test was performed to assess heterogeneity and ensure that each group of studies was suitable for meta-analysis. ORs were pooled using the methods described by DerSimonian,24 and 95% CIs were constructed using Woolf's method.25 If the result of the heterogeneity test was P>0.05, the ORs were pooled according to the fixed-effects model (Mantel−Haenszel methods); otherwise, the random-effects model was used. Funnel plots of the studies' effect estimates against sample size were used to validate the meta-analysis.26 The above statistical analyses were performed using STATA 12.0 . The criterion for statistical significance was set at α=0.05, and all values were two-tailed.Hardy−Weinberg equilibrium testing and allele and genotype frequency analysis were conducted using SHEsis . The frequency of the G allele of rs11098403 was significantly lower among schizophrenia patients than among the controls . There were no significant differences in genotype or allele distributions of the tagSNPs within NDST3 between the schizophrenia and control groups. Analysis of pairwise LD was performed in SNP pairs. We observed two strong pairwise LD blocks in the schizophrenia samples  being dropped. We found no significant association between the haplotypes and schizophrenia. In a comparison between the BD and control groups, no significant difference was observed in individual SNP marker genotypes or allele distributions . We subsequently performed a meta-analysis using the fixed-effects method. After pooling all data from the 1861 patients with schizophrenia and the 2081 controls, we observed a significant association of the rs11098403 G allele with schizophrenia , with an OR=0.70 (95% CI: 0.61−0.79)  . There wNDST3 expression in the brain. There were 19 exon-level and 1 gene-level NDST3 transcripts in the BRAINEAC database. As shown in NDST3 expression in the hippocampus (P=0.027), but this significance did not survive after Bonferroni correction. Compared with the A allele of rs11098403, the G allele showed a higher level of NDST3 expression in the hippocampus.Then, we performed an eQTL analysis to investigate the functional effect of rs11098403 on NDST3 was implicated as a major factor in both schizophrenia and BD pathogenesis; this finding was supported by a GWAS analysis. In this study, we performed a comprehensive investigation of the association of NDST3 with schizophrenia and BD in the Han Chinese. Both our initial results and subsequent meta-analysis replicated the earlier GWAS showing a genetic association between the SNP rs11098403 with schizophrenia in the Han Chinese.Recently, P=6.55 × 109; ref. P=0.03).27 The above findings implied that the rs11098403 G allele was likely to be a protective factor for schizophrenia in the Han Chinese and a risk factor for schizophrenia in Caucasians. To clarify the effect of rs11098403 on NDST3 expression in the brain, we performed an eQTL analysis and observed that rs11098403 might have a regulatory effect on NDST3 expression in the hippocampus; additionally, we found that the G allele resulted in a higher level of NDST3 expression than the A allele. The SNP rs11098403 is located in an intragenic region in the vicinity of NDST3, although our data showed that there was no LD between rs11098403 and the nearby loci within NDST3. NDST3 is an enzyme that is expressed in both the fetal and adult brain, with the highest abundance in the cerebellum and hippocampus.8 This enzyme has a critical role in maintaining heparan sulfate metabolism, and knockout of NDST3 in mice leads to a substantial reduction of heparan sulfate sulfation and affects the physiological functions in the brain.28 There is a growing body of evidence showing that the hippocampus is one of the most significant areas in the pathophysiology of schizophrenia.29 Structural and functional abnormalities in the hippocampus have been frequently reported in patients with schizophrenia. A recent large-scale magnetic resonance imaging data analysis showed that patients with schizophrenia had small hippocampi compared with healthy controls and that hippocampal volume deficits were significantly associated with a high proportion of unmedicated patients.30 The aberrant hippocampal volume could be partially ameliorated by atypical antipsychotic treatment.31 Allen et al.32 measured adult hippocampal neurogenesis in patients with schizophrenia and observed that the cell proliferation marker Ki67 was significantly reduced in their postmortem hippocampal tissues. Taken together, these results indicate that the G allele of rs11098403 is associated with a higher level of NDST3 expression, allowing NDST3 to maintain the normal neurodevelopmental processes in the hippocampus and thereby have a protective role in the development of schizophrenia.In our control subjects, the frequency of the rs11098403 G allele was 17.7%, which was similar to the frequency in the Han Chinese from the HapMap database (frequency=17.4%). Our meta-analysis showed that the OR of the rs11098403 G allele for schizophrenia in the Han Chinese was 0.70 and the 95% CI was from 0.61 to 0.79. However, we noted that the significance of the rs11098403 G allele in schizophrenia appeared in Caucasian populations with OR=1.41  were not included because not enough information was available to reliably include these results in our larger analysis.Several limitations in this study should be noted. First, our eQTL analysis was performed in the BRAINEAC database derived from Caucasian brain samples. It is unclear whether the modulatory effect of the rs11098403 polymorphism affects hippocampal NDST3 expression in the Han Chinese. These results would help to explain the pathophysiological mechanism of schizophrenia.In conclusion, we performed a comprehensive analysis to replicate the early GWAS results in Chinese Han populations. Our findings provided preliminary evidence that rs11098403 might modify the genetic risk of schizophrenia in the Han Chinese. Further investigations are warranted to validate our findings and identify the precise mechanism by which rs11098403 regulates brain


CTLA-4 gene polymorphisms significantly increased the risk of childhood T1D in an allelic model  and a codominant model . Subgroup analysis results indicated that the ORs were higher in the Asian population  than the Caucasian population . The TSA results indicated that the evidence of the effect was sufficient. In conclusion, CTLA4 +49A/G polymorphisms increased the risk of T1D in children, and CTLA4 +49A/G can be considered to be a genetic marker for T1D in children.Type 1 diabetes (T1D) is a heritable disease associated with multiple genetic variants. This systematic review and meta-analysis assessed the correlation between cytotoxic T-lymphocyte-associated protein 4(CTLA-4) +49A/G polymorphisms and the risk of T1D in children. The random effects model was used to estimate the related odds ratios (ORs) and 95% confidence intervals (CIs). Trial sequential analysis (TSA) was used to determine whether the currently available evidence was sufficient and conclusive. Our results indicated that  The cause of T1D is not fully understood. In 1974, Nerup et al. [Type 1 diabetes (T1D) is the most common metabolic disorder in children, and it has serious short-term and long-term implications , and morp et al.  first idp et al. . These sCTLA-4 gene is known to contain multiple single nucleotide polymorphisms (SNPs): -318C/T in the promoter, +49A/G in exon 1, and a multiallelic dinucleotide repeat in the 3′untranslated region (UTR) of exon 4 [CTLA-4 +49A/G locus (rs231775) has a gene dosage effect on the levels of sCTLA-4 [CTLA-4 +49A/G polymorphism has been widely studied in autoimmune diseases, and it has been reported to be involved in the development of Graves’ disease (GD) [CTLA-4 +49A/G polymorphisms in 1996, but this was not further confirmed by Yanagawa T et al.[CTLA-4 +49A/G polymorphisms and the risk of T1D in children is still not confirmed. Since children are a group that is relatively unaffected by confounders , the association between CTLA-4 +49A/G polymorphisms and the risk of T1D in children can more realistically reflect genetic susceptibility to T1D.T1D is an organ-specific autoimmune disease that results from the T cell-mediated destruction of pancreatic β cells . The CTLf exon 4 . The SNPase (GD) , and diaase (GD) . Nisticoase (GD)  first fo T et al. in 1999. T et al.–17 did nCTLA-4 +49A/G polymorphisms and the risk of T1D in children. Furthermore, trial sequential analysis (TSA) was used to determine whether the currently available evidence was sufficient and conclusive.Although Luo et al.  conducteFigure Table As described in Table 2=23.7, P<0.001 for allele frequency and χ2=22.7, P<0.001 for genotype frequency).To assess the allele frequencies in different ethnicities, the allele frequencies in the controls were calculated based on the original data presented in the included studies. The G and A allele frequencies were 43.7% and 56.3%, respectively, in the Asian population, and 37.8% and 62.2%, respectively, in the Caucasian population. The GG, GA, and AA frequencies were 21.5%, 44.5%, and 34.1%, respectively, in the Asian population and 15.9%, 43.8%, and 40.3%, respectively, in the Caucasian population. A statistically significant difference was found between the two ethnicities ; therefore, a random effects model was used and the overall OR was 1.33 (95%CI=1.19-1.48). After the model was stratified by ethnicity, heterogeneity decreased significantly in the Asian population (I2=24.7 and P=0.249), but not in the Caucasian population (I2=53.4 and P=0.012). The ORs were 1.60 (95%CI=1.39-1.83) for the Asian population and 1.24 (95%CI=1.10-1.40) for the Caucasian population. These results did not significantly change after excluding the two studies that did not use HWE (Table Table 1), GA vs. AA (OR2), and GG vs. GA (OR3) were 1.75, 1.26, and 1.35, respectively  and OR2 (2I=40.4% and P=0.036). Thus, a random effects model was applied, and the overall gene effect was significant with a pooled OR1 and OR2 of 1.75  were higher than the ORs for the Caucasian population . The OR value decreased in the age-matched subgroup in comparison to the pooled results.The results of the stratified analyses were not significantly different from the overall results Table . The ORsAn exploration of the sources of heterogeneity was performed using a subgroup analysis stratified by ethnicity, genotyping method, and HWE. As shown in Table Two studies that did not apply the HWE principle were first excluded in the sensitivity analysis, and this did not change the significance of all the ORs Table . Two stuEgger’sPtest>0.05).Figure CTLA4 +49A/G had a 1.33-fold increased risk of developing T1D in comparison to children carrying the A allele; children who had a GG or GA genotype had a 1.75- or 1.26-fold higher risk, respectively, of developing T1D than children who had the AA genotype. These risk effects were further confirmed by TSA.In this study, we found that children carrying the G allele in HLA-DQB1 *02:01 and *03:02 and the haplotypes DR3 and DR4 were significant risk factors for T1D. In contrast, HLA-DQA1*01:01, DQB1*05:03, *06:02, *06:03, and *06:04 were protective factors against T1D. It has been found that HLA-DQ- and HLA-DRB1-containing genotypes contribute to about 50% of the risk associated with developing T1D [PTPN22), and interleukin 2 receptor alpha (IL2RA), are also candidate genes that are associated with the risk of T1D [CTLA4 +49A/G as a genetic marker to assess the T1D risk for children in clinical contexts.Hamzeh et al.  found thping T1D , 37. OthOur results were partly consistent with a meta-analysis performed by Luo et al.  in 2012 CTLA4 +49A/G polymorphism among the controls across different ethnicities. Our study also observed significant difference in the allele and genotypes frequencies in CTLA4 +49A/G polymorphisms between Asian and Caucasian children. Hence, we also conducted a subgroup analysis stratified by ethnicity. The association between CTLA-4 +49A/G polymorphisms and the risk of T1D was greater in Asian children than in Caucasian children, which is consistent with the results of a meta-analysis of all ages [CTLA-4 +49A/G polymorphisms was higher in Caucasians than in Asians. These inconsistent results are understandable because these meta-analyses reviewed different studies with different sample sizes and different inclusion and exclusion criteria. Moreover, other basic characteristics, such as gender and lifestyle, should be considered when explaining the differences.The development of T1D in children has been shown to be caused by different genetic susceptibility factors between different populations . Indeed,all ages . Howeverall ages  reportedCTLA4 +49A/G polymorphisms in different ethnicities. Therefore, because only Asian and Caucasian children were studied in this meta-analysis, further studies on other races are needed.It should be noted that this present study examined significant heterogeneity. Therefore, we performed a subgroup analysis of ethnicity, HWE, and genotyping methods, and the differences between the case studies and the control studies. As expected, heterogeneity decreased significantly in the subgroup stratified by ethnicity, which not only strongly confirms that different races are the main source of heterogeneity in this meta-analysis, but also indicates the different degree of risk of T1D associated with HLA and CTLA4 together; thus, this should be investigated in future studies.Our meta-analysis has several limitations. First, we detected substantial heterogeneity between the examined studies, which would influence the reliability of our conclusions, even if the corresponding pooled OR value was not significantly changed in the sensitivity analysis. Second, we restricted the research to English publications. In addition, one of the included studies focused on West African children. Therefore, potential publication bias should be noted, although a Begg’s funnel plot and an Egger’s test showed no publication bias. Third, the implementation of some of the studies was reported poorly. As shown in Table CTLA4 +49A/G polymorphisms. The TSA results demonstrate that the current evidence is sufficient, thereby suggesting that our results are reliable. Second, this study has more statistical power than previous meta-analyses, and we also selected the best genetic model to avoid multiple comparisons. Third, our study was performed using PRISMA guidelines and the recommendations of the Cochrane Collaboration, although our study was not registered with any specific protocol. In summary, our study is the most recent and the most comprehensive meta-analysis of this topic to date. CTLA4 +49A/G can be considered to be a genetic marker for T1D in children.Despite these limitations, the present study has some advantages in comparison to previous studies. First, we used TSA to determine the association between the risk of childhood T1D and This study was performed according to the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) 2009 statement Suppleme. There wT1DM or T1D or type 1 diabetes or type 1 diabetes mellitus in combination with polymorphism or polymorphisms in combination with CTLA4 or Cytotoxic T-Lymphocyte Associated Antigen-4 or rs231775 in combination with children or child or childhood. In addition, reference lists in the relevant publications were also hand-searched to identify additional records. Only English publications were searched.Two investigators (Bo Wang and Wei Du) conducted a systematic literature search of the EMBASE, PubMed, and Cochrane Library databases for papers related to the studied topic that were published up to March 30, 2016. The following search terms were used: CTLA4 +49A/G polymorphisms and T1D risk in children (≤16 years); moreover, the genotype frequencies of CTLA4 +49A/G loci in the cases and controls could be extracted from these studies (or could be obtained after contacting the authors).Two reviewers (Bo Wang and Wei Du) identified eligible articles based on a full-text review. To be included, studies had to be case-control or cohort studies describing the association between Exclusion criteria were: studies with a sample size <100 for limited statistic power; in addition, family-based studies, review articles, letters, case reports, editorials, and conference abstracts were also excluded. Any disagreement was resolved by consensus.Two investigators (Bo Wang and Wei Du) independently extracted data from the included studies, and it was critical that the final data extracted by the two investigators be the same. Disagreements were resolved by discussion and consensus. The data included: the first author’s name, year of publication, the country and ethnicity of the cases, the age of the patients and healthy controls, the sample size, the diagnosis criteria, the genotyping method, and the genotype frequencies of the cases and controls. Hardy-Weinberg equilibrium (HWE) was calculated from the extracted data. A single article with different populations was further categorized as different studies. If the literature did not provide sufficient data, the investigators tried to contact the author by email to obtain the original data.To assess the quality of each study, we used the widely-used quality assessment criteria for genetic association studies defined by Thakkinstian et al. , with soCTLA4 gene +49A/G polymorphisms and T1D risk in children. Previous studies have indicated that G is the risk allele; therefore, we first calculated the ORs of GG vs. AA (OR1), GA vs. AA (OR2), and GG vs. GA (OR3), and we then selected the best genetic model based on the meta-analysis method used in molecular association studies [In this study, unadjusted odds ratios (ORs) and the corresponding 95% confidence intervals (CIs) were used to measure the strength of the association between  studies .2I) statistic were calculated, and a fixed effects model was used if 2I<50% and P≥0.1, whereas a random effects model was used if 2I≥50% and P<0.1 [To assess for the heterogeneity of the ORs across studies, the Cochrane Q statistic and I-squared  and RevMan 5.3 (Cochrane Collaboration).http://www.ctu.dk/tsa) was used in this study. When the cumulative Z-curve crosses the trial sequential monitoring boundary or enters the futility area, a sufficient level of evidence may have been reached, and no further trials are needed. If the cumulative Z-curve does not cross any of the boundaries, and the required information size has not been reached, there is insufficient evidence to reach a conclusion [Meta-analyses may result in type I errors due to random error from the studies included in the meta-analysis that had a small sample size, publication bias, and low quality, and studies whose conclusions tended to be changed by later studies with a larger sample size . TSA cannclusion .


Single nucleotide polymorphisms in microRNAs  have been inconsistently associated with risk for oral squamous cell cancer (OSCC). This meta-analysis aimed to assess the correlation between microRNA polymorphisms and susceptibility to OSCC.Free words were used to search for the relevant studies without language limitations in electronic databases including PubMed, Embase, Web of Science and SCOPUS through June 15, 2017. Odds ratios (ORs) were calculated to investigate the effects of microRNA polymorphisms on oral cancer risk.P < 0.05). MicroRNA-196a2 rs11614913 and microRNA-149 polymorphisms appeared to have no relationship with OSCC risk (P > 0.05). In the sensitivity analysis, there was a significant association between the TT genotype of microRNA-196a2 and OSCC risk .Eleven studies were included. Analysis under the recessive model of microRNA-146a (CC vs GG+CG) showed significant differences . The G allele and the GG genotype of microRNA-499 were associated with OSCC risk (ORs >1, There may be no significant relationship between microRNA-149 polymorphisms and OSCC risk, and the CC genotype of microRNA-146a may have protective effects against oral cancer. However, the G allele and the GG genotype of microRNA-499 may increase OSCC risk. The Incidence of oral cancer cases has been steadily increasing, due to population growth and changes in age structure . GloballMicroRNAs (miRNAs), which are endogenous, small non-coding RNAs, are critical to the regulation of physiology and disease progression, including oral cancers . MiRNAs MiRNAs exhibit remarkable tissue specificity , and miRAbnormal microRNA polymorphisms in OSCC warrant more research to provide a basis for diagnosis, therapy and screening for the high-risk OSCC population. Advances in our efforts to detect the role of miRNAs in genetic susceptibility to OSCC may further elucidate their role in regulating OSCC risk and may identify certain miRNAs as oncogenic biomarkers. Therefore, the purpose of this study was to review the literature for evidence of correlations between certain microRNA polymorphisms and the risk of oral squamous cell cancer.The initial electronic database searches of PubMed, Embase, Web of science and SCOPUS yielded 967, 18, 160, and 110 potentially studies, respectively. After reading the titles and abstracts of those studies, 20 articles were chosen for full-text reading, and their references were screened. Finally, 11 eligible studies on SNPs in miRNAs were included after the unqualified studies were discarded . The ORs in all 5 models were close to 1, which may suggest that the microRNA-196a2 rs11614913 polymorphism has a very low effect on OSCC risk. However, the sensitivity analysis, which was conducted by eliminating one study at a time [rs11614913 decreased OSCC risk   . Heterogeneity was not found in any of the genetic models . The high level of agreement among the included studies provided powerful evidence that there is a strong negative correlation between the CC genotype of microRNA-146a rs2910164 and OSCC risk.A pooled analysis of 6 studies , 17, 18 ) Figure . The gen2 ≥ 50%). Therefore, further well-designed studies investigating the relationship between the microRNA-149 rs2292832 polymorphism and risk of OSCC are warranted.For microRNA-149 rs2292832, 3 studies , 12, 15 2 = 39.9%, OR = 1.57, 95% CI = 1.318∼1.871, P = 0.000), the additive model , the codominant model , the dominant model  and the recessive model . The high ORs in all genetic models (ORs > 1) indicate that the G allele and the GG genotype of microRNA-499 rs3746444 increase the risk for oral cancer.As indicated in Table  when three studies were excluded [To confirm the outcome of our analyses, we conducted a sensitivity analysis by sequentially eliminating one study at a time. The sensitivity analysis revealed that certain studies significantly influenced the correlation between microRNA polymorphism and risk of OSCC. The TT genotype of microRNA-196a2 rs11614913 was found to potentially decrease OSCC risk excluded , 17 inflEgger> 0.05, PBegg> 0.05) using Stata version 14.0 software . No significant publication bias was observed, except in the case of the additive model of microRNA-499  genes, miR-196a plays an important role in regulating HOX expression and function . By targFor microRNA-146a, the G > C substitution within the pre-miR-146a sequence resulted in less conversion into mature miR-146a by affecting the efficiency of pri-miR-146a processing. Additionally, that substitution disturbed protein binding to pre-miR-146a . IncontrThe miR-149 polymorphism involves a C>T substitution. Through diverse signaling pathways, miR-149 targets Akt1, a gene that inhibits tumor cell apoptosis, thereby exerting its oncogenic potential . No assoLocated in the 3p mature miRNA, the miR-499 rs3746444 A/G polymorphism interferes with miRNA binding and pre-miRNA maturation . AlthougMany limitations might have influenced our findings. The most obvious limitation was the small number of included studies. Further research investigating microRNA polymorphisms and oral squamous cell cancer risk is warranted to confirm these findings. Furthermore, the poor consistency indicated by the sensitivity analysis of the recessive models of microRNA-196a2 and microRNA-146a might have influenced the statistical power of our study. Selection bias of included controls should be avoided by basing selection on the population rather than the hospital. Moreover, environmental factors, such as habits, should be balanced among studies. Other limitations of this study include a lack of subgroup analysis, disparity in OSCC location and the fact that we did not clarify non-response rates. Our meta-analysis failed to avoid these limitations due to a deficiency of data.To date, studies concerned with microRNA polymorphisms and the risk of oral squamous cell cancer are insufficient and inconsistent. This review focused on four microRNA polymorphisms. The results of our meta-analysis suggest that the microRNA-196a2 and microRNA-146a polymorphisms may increase OSCC risk. In contrast, the miR-499 polymorphism may exert protective effects against OSCC risk. There may be no significant association between the microRNA-149 polymorphism and OSCC risk.Due to the small number of included studies, the associations between the selected SNPs of these four microRNAs and OSCC risk are weak.The protocol details were registered with the public registry of systematic reviews, PROSPERO (CRD42017073480). Our study was conducted by following the recommended Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement. A search of PubMed, Embase, Web of Science and SCOPUS was conducted through June 15, 2017, to identify all the relevant studies without language limitation. The search strategy, which employed free-text words, is shown in Table Next, titles and abstracts were read manually for preliminary analysis, in duplicate, by two reviewers. Disagreement was resolved by discussion or by referring the debate to a third reviewer. Finally, the chosen articles were subjected to the following inclusion and exclusion criteria:1. The included articles should be case-controls. 2. Patients with a definite diagnosis of oral cancer, not oral precancer, should be mentioned. 3. The included studies must contain extractable data on SNPs in pre-miR distribution in cases and controls.P value for HWE (PHWE) in the control group was less than 0.05 would be eliminated from the sensitivity analysis. 5. Studies are repeated studies by the same author or team.1. Articles are review articles or they describe experiments in animals. 2. Studies focus on miRNA-processing genes or miRNA-binding site genes instead of pre-miRNA genes. 3. Studies report non-extractable data. 4. Studies in which the http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp) independently and strictly.Each chosen study was subjected to thorough reading for data extraction, synthesis, analysis and quality assessment, following the Newcastle-Ottawa Scale (NOS) system  was confirmed in PubMed. The outcome variables consisted of the allele and genotype frequencies of miRNA polymorphisms in cases and controls. Stata version 14.0 software was used for the data analysis. The odds ratio (OR) and 95% confidence interval were calculated for five genetic models: the allele model, the additive model, the codominant model, the dominant model and the recessive model. The


A number of genetic loci were found to be associated with dystonia. Quite a few studies have been contacted to examine possible contribution of TOR1A variants to the risk of dystonia, but their results remain conflicting. The aim of the present study was to systematically evaluate the effect of TOR1A gene SNPs on dystonia and its phenotypic subtypes regarding the body distribution.G) model were used to calculate both the pooled point estimate in each study and the overall estimates.We performed a systematic review of Pubmed database to identify all available studies that reported genotype frequencies of TOR1A SNPs in dystonia. In total 16 studies were included in the quantitative analysis. Odds ratios (ORs) were calculated in each study to estimate the influence of TOR1A SNPs genotypes on the risk of dystonia. The fixed-effects model and the random effects model, in case of high heterogeneity, for recessive and dominant mode of inheritance as well as the free generalized odds ratio (ORG (95%C.I.): 2.58 (1.45–4.58)].Rs1182 was found to be associated with focal dystonia in recessive mode of inheritance . In addition, rs1801968 was associated with writer’s cramp in both recessive and dominant modes  and [2.48 (1.36–4.51), Pz = 0.003) respectively and in model free-approach [OROur meta-analysis revealed a significant implication of rs1182 and rs1801968 TOR1A variants in the development of focal dystonia and writer’s cramp respectively. TOR1A gene variants seem to be implicated in dystonia phenotype. Dystonia is a common but heterogeneous movement disorder. It is estimated to be the third most frequent movement disorder worldwide . HoweverRecently, a new general definition and a new classification of dystonia have been proposed . ClassifTOR1A gene  covers an 11k bp region in chr9 and it is consisted of 5 exons. TOR1A protein, called TorsinA belongs to the family of the AAA+ ATPases that can be found in the endoplasmic reticulum and nuclear envelope of most cells  includinRelatively few genetic loci have been identified as potential causing factors of hereditary forms of isolated and combined dystonia, with a wide variation in the mode of inheritance, clinical features, body distribution and age of onset. The DYT1 (TOR1A) form of hereditary dystonia is mainly a generalized  dystonia, which is inherited by an autosomal dominant mode . RegardiTwo meta-analyses evaluated, so far, the effects of TOR1A gene variants on primary dystonia , 24, wheThe previous two meta-analyses , 24 usedS1 Appendix. We imposed no language or other restrictions. Last literature search was performed on September 9th, 2016. The reference lists of all retrieved articles were additionally examined to identify studies that may have been missed by the initial database search. The inclusion criteria were a) case-control studies, in which cases were clinically diagnosed of having dystonia and controls were neurologically healthy, b) studies reporting single nucleotide polymorphisms in the TOR1A gene and c) studies that mentioned in the text the genotype frequencies from patients and controls. Studies with incomplete data or without genotype frequencies were excluded from meta-analysis. The initial phenotypic classifications of participants in each study were maintained.Eligible case-control candidate gene association studies (GAS) were selected by searching Pubmed database. The combination of search strings that was used included the following terms: “dystonia” and “tor1a” and “polymorphism”. The complete search algorithm is available in the The following information was extracted from each study: author, year of publication, ethnicity of the studied population, numbers of cases and controls, age at disease onset, mean age and gender distribution, genotype frequencies, tested polymorphisms, family history of the participants, method of diagnosis, screening or not of TOR1A ΔGAG mutation, deviation or not from the Hardy-Weinberg Equilibrium (HWE), and tested dystonia phenotypes.Fig 1.All individual studies were reviewed and the risk of possible bias was assessed. The flowchart presenting the selection procedure of eligible studies is presented in G) and the corresponding 95%C.I. were applied in order to quantify the association between genotype distribution and disease. ORG is an inheritance model free approach that estimates the mutational load in cases compared to the controls  and the recessive [(mt/mt) vs (wt/mt+wt/wt)] genetic modes of inheritance. The significance of the OR was determined by the Z test . Additionally, the generalized odds ratio . Otherwise the fixed-effects model  was appl method ) was useTest for possible publication bias was graphically assessed using the funnel plot. Furthermore, it was evaluated by the linear regression asymmetry test by Egger , when ithttp://tech.cochrane.org/revman)]. The ORG was calculated with ORGGASMA (www.biomath.uth.gr) software. The PRISMA guidelines for reporting reviews and meta-analyses (S2 Appendix) were applied in this meta-analysis.All statistical analyses were performed in Review Manager (RevMan) Version 5.2 software . The main meta-analysis results for the entire focal dystonia group and the focal dystonia subgroups  are shown in Fig 5. The respectively results after the meta-analysis with the model-fee approach and the genotypes frequencies, are showed at Table 1for the overall dystonia group and at Table 2for the focal dystonia group and focal dystonias’ subtypes. Overall, rs1182 was found to be associated with focal dystonia in recessive mode . Moreover, rs1801968 has found to be associated with writer’s cramp in both recessive and dominant modes  and  respectively and in model free-approach [ORG (95%C.I.): 2.58 (1.45–4.58)].The main results of meta-analysis for all available SNPs in the overall dystonia group are presented as forest plots in In the present meta-analysis, that included a relatively large number of participants, we investigated the effect of TOR1A gene SNPs on the risk of dystonia, as well as on the risk of focal dystonia and its subtypes . Our study detected a significant influence of a specific variant of TOR1A gene, rs1182, on the risk of focal dystonia. It also showed and that there is a strong association between rs1801968 and the development of task specific writer’s cramp focal dystonia. To the best of our knowledge, this is the first meta-analysis examining the effect of TOR1A gene polymorphisms on the above-mentioned disorders.Quite a few polymorphisms across TOR1A gene have been examined for possible association with dystonia, but the results from candidate gene association studies (CGASs) remain conflicting , 23. AddThe ΔGAG human variant of TorsinA was the first pathogenic, disease-causing mutation identified in early onset, generalized dystonia . ThereafRs1182, that also reached significance in our study, is located in the 3’ untranslated region (3’-UTR) of exon 5 of TOR1A gene. However, there is no known functional effect of rs1182 on expression and function of TOR1A  so, the 2: 0–28%). However, it is not known whether these associations are causal without supportive functional analyses.Both variants emerged in our analysis may have some functional consequences. In particular, rs1182 is located in the UTR-3 that may affect transcription of TOR1A gene, whereas rs1801968 represents a missense mutation and is characterized as pathogenic in NCBI. In addition, an almost consistent effect across varied populations is also supportive of the significant role of both polymorphisms and by extension TOR1A gene. Moreover, the heterogeneity of effect size across studies it remained within acceptable limits Click here for additional data file.S2 Appendix(PDF)Click here for additional data file.S3 Appendix(PDF)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.


I2 = 73%, p = 0.000; milk (6 studies): 0.99, 95% CI: 0.89–1.10, I2 = 0.0%, p = 0.43; cheese (5 studies): 0.89, 95% CI: 0.76–1.05, I2 = 39%, p = 0.16). The only cohort study with a total of 456,513 participants reported a positive association of butter intake with endometrial cancer risk . There was a significant negative association of dairy products intake and endometrial cancer risk among women with a higher body mass index (BMI) . Stratifying the analyses by risk factors including BMI should be taken into account when exploring the association of dairy products intake with endometrial cancer risk. Further well-designed studies are needed.Observational studies have suggested inconsistent findings on the relationship between dairy products intake and endometrial cancer risk. This study aimed to conduct a meta-analysis to evaluate this correlation; moreover, databases including PubMed, ISI Web of Science, and Embase were screened for relevant studies up to 26 February 2017. The inverse variance weighting method and random effects models were used to calculate the overall OR (odds ratio) values and 95% confidence interval (CI). A total of 2 cohort study and 16 case-control studies were included in the current analysis. No significant association was observed between endometrial cancer risk and the intake of total dairy products, milk, or cheese for the highest versus the lowest exposure category : OR 1.04, 95% CI: 0.97–1.11,  According to the GLOBOCAN 2012, endometrial cancer is the fifth most common cancer in women over the world  OR diets [tiab] OR dietetic [tiab] OR dietary [tiab] OR eating [tiab] OR intake [tiab] OR nutrient* [tiab] OR nutrition. The specific keywords about dairy products included “dairy [tiab]” OR “milk [tiab]” OR “yogurt [tiab]” OR “cheese [tiab]”. For the outcome terms, they were consistent with terms published in a previous meta-analysis [According to the WCRF Specification Manual  the study design was a case-control study or a prospective study; (2) the exposure included dairy products, milk, yogurt, butter, and cheese; (3) the outcome was endometrial cancer; (4) the odds ratio (OR) or relative risk between the intake of dairy products and the risk of endometrial cancer was reported. We identified relevant 1023 articles from our search on PubMed, 1286 studies from the ISI Web of Science, with literature of Editorial, case report, review, book, news, reference material, meeting, clinical trial, patent, biography, letter excluded, and 1129 relevant articles from our search on Embase. Titles and abstracts were reviewed independently, and fifty pertinent articles were retrieved . After gData extraction forms for recording study characteristics, quality issues, and results of studies were developed to get as much details of the studies included as possible. Two investigators  independently extracted the details of the study design, including population, exposure, outcome assessment, sample size, participant characteristics , menopause status), time frame, survey method, and statistical analyses, as well as confounders for adjustment from the included studies . The NewAccording to the WCRF criteria, the exposures evaluated in more than two cohorts or five case-control studies were used for analysis. There was a sufficient number of studies to conduct a meta-analysis for total dairy and milk intake. Seven studies ,25,26,29p-value was less than 0.05 in the Q test, or the I-squared was more than 50%, heterogeneity was considered statistically significant, and the random effect model was used for further analysis. Moreover, to search for potential sources of heterogeneity, subgroup analyses were performed by geographic region, age, BMI, and menopausal status. Concerning the lack of the age- and BMI-based group-specific data on endometrial cancer risk, the median or mean value of age and BMI among controls were used to divide the studies to two relative groups. The publication bias was evaluated with funnel plots and the Egger’s regression asymmetry test, i.e., if the p-value was less than 0.05, this indicated a significant publication bias. Furthermore, to investigate the impacts of individual studies on the overall results, we also performed a sensitivity analysis by recalculating the overall OR value after removing each study one at a time. The heterogeneity across the studies was estimated using the Mantel–Haenszel model, i.e., if the All statistical analyses were performed with STATA version 12.0 . All statistical tests were two-sided. For the current meta-analysis, we included two cohort studies which included two populations, seven population-based case-control studies, eight hospital-based case-control studies, and one pooled analysis of three Italian hospital-based case-control studies. The population in the study conducted by Filomeno et al.  includedI2 = 73%, p = 0.000) between total dairy intake and endometrial cancer risk when comparing the highest category of total dairy intake to the lowest category. Based on the 13 case-control studies, the estimated OR was 0.87 , and the OR for the cohort studies was 1.04  (p = 0.0499) . After excluding the studies with NOS stars less than 6, the OR for case-control studies was 1.05 , and the OR for cohort studies was 1.05   . Funnel  0.0499) , reflect = 0.37) . PoolingI2 = 80.2%, p = 0.000) in North America, 1.01  in Europe, and 1.01  in Asia. The analysis stratified by BMI showed an OR of 0.66  in the group where the median BMI was greater than 25, and 1.05  in the group where the median BMI was no more than 25. Additionally, in the subgroup classified by median age, the group older than 55 showed an OR for risk of endometrial cancer of 0.86  when comparing the highest category with the lowest category of total dairy intake, whereas the group younger than 55 showed an OR of 1.03 . When stratified by menopause status (menopause frequency greater or less than 70%), the ORs of dairy intake and endometrial cancer risk were 0.86  and 1.06 , respectively.With the limited amount of studies, we conducted a further stratified analysis using the following parameters: geographic region, BMI, age, and state of menopause. In the analysis stratified by geographical region, the OR for total dairy intake and risk of endometrial cancer was 0.85 . The overall OR was 0.99   .I2 = 59.5%, p = 0.09)  was observed also for the two cohort studies.The pooled estimate of three case-control studies ,19,26 sh = 0.09) . No signI2 = 2.6%, p = 0.31) (Only the two cohort studies  tested t = 0.31) .In this meta-analysis, no significant association between total dairy products intake and endometrial cancer risk was observed. Though geographic region, BMI, age, and state of menopause did not modify the association significantly, we observed a relatively lower risk of endometrial cancer associated with dairy product intake in women with a higher BMI , and thiThe exposure to dairy products was not considered in the Endometrial Cancer 2013 Report and its effect remained limited with no conclusions, based on the meta-analysis conducted in 2007 . CompareDairy products contain saturated fatty acids and estrogen which were reported to be positively associated with endometrial cancer risk ,36. HighIndeed, the ingredients in whole milk, low-fat milk, cheese, yogurt, and butter differ a lot. Butter contains more fat while cheese contains more calcium. When we explore whether dairy products intake is protective or harmful for endometrial cancer, we need to consider the varying ingredient content in different dairy products. Furthermore, no significant association was found between endometrial cancer risk and milk intake, as well as cheese intake in the current study. Only the two cohort studies  reportedThis meta-analysis has several strengths. The sensitivity analyses were conducted to figure out the influence of each study on the summary of the estimates. This is the first time that the potential effects of modifications of the dairy intake on endometrial cancer risk are evaluated by geographic region, age, BMI, and menopausal status. Limitations of this meta-analysis should be noted too. Firstly, the number of studies eligible for the meta-analysis was limited, and 16 out of 18 reports were case-control studies, which were susceptible to recall a bias. In addition, the sample size in six case-control studies ,24,31 waIn summary, this meta-analysis did not observe a statistically significant association between overall dairy products intake and endometrial cancer risk. However, increased cancer risk was suggestively associated with butter intake, even if this result was based on only two studies. There was a significant negative association of dairy products intake and endometrial cancer risk among women with a higher BMI. Future well-designed prospective studies with precise measurements of dairy products exposure with biomarkers, data from various continents, and stratified analyses by risk factors including BMI are needed to evaluate the effect of different types of dairy products on endometrial cancer risk.


We aimed to elucidate the association between fat mass and obesity associated gene (FTO) polymorphism and the risk of polycystic ovary syndrome (PCOS) by meta-analysis.2I and Q statistics. 2I>50% or P<0.05 was considered as heterogeneity statistically, and random effects model was used for pooled analysis. Otherwise, fixed-effect model was used.We searched PubMed and Embase databases to find the relevant studies. Odds ratios (ORs) and their corresponding 95% confidence intervals (CIs) were used for pooled analysis. Statistical analyses were carried out by using R 3.12 software. Heterogeneity was assessed using Twelve eligible studies that published from 2008 to 2015 were included in this meta-analysis. The pooled analyses showed that rs9939609 polymorphism of FTO gene was significantly associated with risk of PCOS under A vs. T, AT vs. TT, AA vs. TT, AA vs. AT+TT and AA+AT vs. TT genetic models. However, for rs8050136 and rs1421085, significant association was only found under recessive genetic model.rs9939609 variation of FTO gene is significantly associated with risk of PCOS. However, the association between rs8050136, rs1421085, and PCOS is still unclear and needs further confirmation. Polycystic ovary syndrome (PCOS) is a common endocrine disease among women of reproductive age –3. The tUp to now, the treatment of PCOS is still a challenge for physicians and medical researchers. The treatments are mainly adopted depending on the symptoms of the PCOS patients. For examples, letrozole and clomiphene are used to treat the infertility of PCOS . Some inBecause of the difficult treatment of PCOS, the medical researchers have tried to consider the risk factors associated with PCOS to prevent prevalence of the disease. Genetic and environmental factors have been identified to be associated with PCOS. The fat mass and obesity-associated gene (FTO) as an obesity candidate gene has been related to PCOS susceptibility . HoweverIn this study we systematically meta-analyzed the genotypes of FTO gene and risk of PCOS, thus, to identify key variations of FTO gene associated with risk of PCOS.We searched two English databases PubMed and Embase using the search terms of (PCOS OR polycystic ovarian syndrome OR Polycystic ovary syndrome) and (FTO OR fat mass and obesity associated gene OR s1121980 OR rs1421085 OR rs1558902 OR rs8050136 or rs17817449 or rs9939609 or rs9930506). The deadline of the search was Mar 2016. Web of science and Google scholar were also used for checking leakage.Inclusion criteria included: 1) the study reported the association between FTO and FCOS; 2) the study provided the data for calculating the distribution of FTO in PCOS patients and non-PCOS population. The reviews, reports, comments and letters were excluded from this analysis.The following data extracted from included studies: name of the first author, publication year, location of the study, the number of PCOS patients (cases) and non-PCOS population (controls), age of the participants and Body Mass Index (BMI). The number of each genotype in cases and controls was also extracted.Quality assessment was conducted using Newcastle-Ottawa Scale (NOS)  independ2I and Q statistics  test was performed using chi-square . Pooled atistics . I2>50% ion bias .The process of study selection was shown in Totally, 63 studies were identified by searching PubMed and Embase, and no additional studies were found from Web of science and Google scholar. Firstly, 17 duplicate studies were removed. Secondly, 19 irrelevant studies were excluded after reading titles and abstracts. Thirdly, 14 studies were removed from remaining 26 studies after reviewing full-texts. Finally, 12 eligible studies , 15–23 wP<0.05).The characteristics of 12 case-control studies were shown in 2I=61.2%), additive , recessive  and dominant genetic models . Therefore, the random effects model was used for pooling analysis. The analyses under other genetic models were pooled using fixed effect model.We meta-analyzed the association between PCOS and FTO gene polymorphisms at loci of rs9939609 (A vs T), rs8050136 (A vs C) and rs1421085 (C vs T) under allele, additive, recessive and dominant genetic models. As shown in The pooled analyses showed that single nucleotide polymorphism (SNP) at rs9939609 of FTO gene was significantly associated with risk of PCOS under A vs. T , AT vs. TT , AA vs. TT , AA vs. AT+TT  and AA+AT vs. TT  genetic models. However, for the SNPs at rs8050136 and rs1421085, significant association was only found under AA vs. AC+CC , which indicated that the pooled results in this meta-analysis were credible.Egger’s test  showed nIn this study, we conducted a meta-analysis on the association between FTO gene polymorphisms and the risk of PCOS. The mutation at rs9939609 loci of FTO could significantly increase the risk of PCOS. However, no reliable evidence can identify the relationship between the other two SNPs of FTO (rs8050136 and rs1421085) and PCOS.Our results are opposite to a previous meta-analysis which states that rs9939609 variant of FTO gene has no association with PCOS . Rs99396Except for obesity and race, metabolic syndrome may be another important factor influencing the association between FTO polymorphism and PCOS. Rs1421085 (C/T) polymorphism in FTO showed a significant association with obese PCOS women or PCOS patients with metabolic syndrome, but not associated with lean PCOS patients or controls . In addiThere are several limitations in this meta-analysis. Firstly, some covariates were not adjusted because of lack of studies and incomplete data. Secondly, the influences of some variates of FTO  on PCOS were not meta-analyzed due to insufficient data. Thirdly, HWE test showed disequilibrium in controls in several included studies. Therefore, these controls may not predict the condition of the population in a certain region. Despite all this, our study can provide an important reference for the further researchers and help them understand the association between FTO polymorphism and PCOS systematically.Rs9939609 SNP variation in FTO gene can significantly increase the risk of PCOS in women. However, rs8050136 and rs1421085 are associated with PCOS under recessive model, but no association under other genetic models. Therefore, the association between these two SNPs and PCOS are still unclear. Further studies with large-scale and high quality are needed to elucidate the association between them. Our study would provide reference for understanding the relationship between FTO and the risk of PCOS and help clinicians take strategy to treat PCOS.Ethical issues  have been completely observed by the authors.


Some controversies still exist between the detection of Epstein–Barr virus (EBV)'s DNA and risks of periodontal diseases. Hence, a comprehensive meta-analysis on all available literatures was performed to clarify the relationship between EBV and preidontitis.A comprehensive search was conducted within the PUBMED, EMBASE, and WANFANG databases up to October 10th, 2016 according to inclusion and exclusion criteria and finally 21 case–control literatures were obtained. The outcomes including odds ratios (ORs) with 95% confidence intervals (CIs) were used to assess the strength of associations. Publication bias was determined by Begg or Egger test. Sensitivity analysis was used to investigate reliability and stability of the results.P < 0.001). In the disease-type analysis, the pooled ORs for chronic periodontitis and aggressive periodontitis were 6.586  and 8.361 , respectively. In the subgroup analysis of ethnicity, our results suggested that high EBV-detecting frequencies were correlated with increased risks of periodontitis in Asians, Europeans, and Americans (P < 0.001). Subgroup analysis by the sample type showed that subgingival plaque (SgP) samples and tissue samples were available for EBV detecting (P < 0.001). Detecting EBV of samples in ≥5 (6) mm sites of periodontal pockets were easier than in ≤3-mm sites (P = 0.023).According to the data from included trials, the association between overall increased risks of periodontitis and the detection of EBV was significant  mm sites of periodontal pockets are more sensitive than in ≤3-mm sites. The subgingival plaque is required for the initiation of the disease. Interestingly, several studies suggested that the current theory of bacterial plaque cannot explain that patients who are absent of these specific bacterial species still got periodontal diseases.,4 And no significant difference in the prevalence of bacteria between healthy and diseased periodontal tissues has been found. Therefore, human herpesviruses have been found to be involved in the etiology of periodontitis because bacterial activity alone is not able to explain all the clinical characteristics of periodontal diseases.Periodontitis is a chronic inflammatory disease that is characterized by periodontal damage, alveolar bone resorption, pain, and eventual tooth loss. The first report about the relationship between EBV and chronic periodontitis (CP) came into our sight in 1996. Afterwards, a number of articles–29 have investigated the associations between EBV and periodontal diseases including CP and aggressive periodontitis (AgP). However, these findings were full of controversy among detecting EBV existence in the periodontal environment. Some studies–23,26–29 have reported that with high prevalence of EBV DNA detecting, the risks of periodontal diseases are significantly increased; whereas others,16,24,25 suggested a weak or even no relationship between them. Hence, we performed the current comprehensive meta-analysis, which combines results from literatures to confirm whether the EBV is associated with periodontal diseases.EBV, also called human herpesviruses 4 (HHV-4), belongs to γ-herpes virus subfamilies. EBV has widely infected >90% adults in the world and is associated with many human diseases, such as post-transplate lymphoproliferative diseases, nasopharyngeal carcinoma, and oral hairy leukoplakia.22.1A systematic search was conducted within PubMed, Embase, and Wanfang databases up to October 10th, 2016 by ZG and MW, using the following key terms: “EBV” and “periodontitis OR periodontal disease.” Reference lists of selected literatures were examined manually for other available publications.2.2The studies were included with the following inclusion criteria: the experimental design was a case–control study; the topic estimated the association between periodontal diseases and EBV; (iii) sample-extracting methods were limited as following: surgery, paper point, curette, paper strip, and biosy; (vi) the patients and controls must be systemically healthy; (v) articles must offer the sample sizes, odds ratios (ORs), and its 95% confidence intervals (CI) or sufficient data to evaluate the association between periodontal diseases and EBV. The major exclusion criteria were as follows: no case–control studies; participants with systemic disease; saliva samples; (vi) no useful data could be extracted or obtained; (v) diseases are not diagnosed as periodontitis or periodontitis-similar diseases.2.3The following information was selected independently by 2 authors (ZG and JL) according to the criteria listed previously: the first author's name, publication year, country, sample size, sample type, sampling method, and disease type. All controversial questions were resolved by asking a third author. All extracted data were based on previous published studies; thus, no ethical approval and patient consent are required.2.4P value of I2 was <50%; otherwise, the random-effects model was chosen. ORs and the corresponding 95% CIs were conducted to evaluate the association between detection of EBV and periodontal disease risks. Sensitivity analysis was investigated to assess the robust of pooled results by omitting one study each time. The publication bias was determined by the Begg rank correlation test and Egger linear regression test. P < 0.05 was considered statistically significant, and all P values were 2-sided. The ORs and 95% CIs in this meta-analysis were performed using Stata 12.0 .First, the heterogeneity test was detected with use value. A fixed-effects model was selected if the 33.1–29 including 995 patients and 564 healthy people, were eventually included in our meta-analysis. Figure Total 710 potential relevant literatures were collected from the initial search. After duplicates removing with EndNote, 142 studies were remained. Through screening titles and abstracts, 79 irrelevant studies were excluded; the remaining 60 records, which investigated the associations between EBV and different periodontitis, were eligibly evaluated with full-text reading. According to our inclusion criteria, 21 studies,Among all the selected studies, 15 were conducted in CP patients and 8 were in AgP patients. Probands of 10 studies were in Asian participants, 8 in whites, and 3 in Americans. Sample types included 19 studies in subgingival plaque (SgP) samples, 2 in tissue samples, and 1 in gingival crevicular fluid (GCF) samples. Three studies reported detecting EBV in different depths of periodontal pocket of periodontitis patients. Table 3.2P < 0.001, I2 = 70.5%); a random-effect model was assumed. The pooled results showed that EBV infection is associated with the increased risk of periodontitis  containing 995 patients with periodontal diseases and 564 periodontal healthy controls were included. The forest plot is shown in Figure 1 contain3.3P < 0.001) and AgP .Subgroup analysis by disease types of periodontitis indicated that high detecting frequencies of EBV were significantly associated with increased risks of CP , Europeans , and Americans  and tissue sample  were associated with periodontitis. However, only 1 study reported that detecting EBV by GCF sample was not associated with CP . After excluding this study by Grenier et al, the meta-analysis of remaining studies become stable . The results of sensitivity analysis for other meta-analysis were remained similar after excluding 1 study each time.One included study of this meta-analysis was omitted each time to evaluate the stability of pooled results. The result of sensitivity analysis indicated that the study published by Grenier et alBegg test and Egger test were used to evaluate the publication bias; the results were summarized in Table 4 Although a variety of presume bacteria are considered to be indispensable in the initiation of periodontitis, it is hard to explain the following clinical characteristics: rapid attachment loss and bone destruction with minimal plaque, special site in periodontitis, and existences of disease activity and quiescence phases.,30 So more and more studies have paid attention to the relationship between herpes viruses and different types of periodontitis.–29 In this study, we focused on whether EBV, a kind of γ-herpes viruses, has an association with increased risks of periodontal diseases.Periodontitis is a chronic disease that affects the majority of adults around the world.–23,26–29 However, there are still some controversies in these findings. Some literatures indicated that a weak or even no relationship exists between herpes viruses and risks of periodontitis.,16,24,25 A meta-analysis has summarized the relationships between risks of CP and herpes viruses. In this study, we made a comprehensive meta-analysis to clarify the associations of EBV and periodontal diseases based on existing research data.Numerous studies have reported that herpes viruses, especially EBV and human cytomegalovirus have significant associations with increased risks of varieties of periodontitis, such as CP and AgP. In addition, we, for the first time, made a meta-analysis to demonstrate that EBV was associated with the increased risk of AgP  mm from periodontal pockets of patients, our result suggested that it was easier to detect EBV in deep sites than shallow sites  before the initiation of periodontitis, and the extent of EBV in pECs is increased with periodontitis severity. Similar results were described by Kato et al. The ORs of co-infection between EBV and Porphyromonas gingivalis were also higher in patients with CP than in health donators.,35 Interestingly, Saygun et al found that the EBV has a correlation with P gingivalis but not A actinomycetemcomitans in AgP. Above all, no matter in CP or in AgP, the co-infection between EBV and P gingivalis may be a kind of fixed group of periodontopathic microorganisms. Studies suggested that EBV infection has an association with a median 1.5 times further increase of the high cytokine TNF-α expression, inflammatory processes of bone destruction and clinic symptomatic expressions in periapical lesions. Other studies included in our meta-analysis only evaluated the relationships between EBV and periodontal diseases.We did not include the studies evaluating the relationship between other herpes viruses and periodontitis because the main purpose of this meta-analysis is to assess the association between detecting of EBV as a single risk factor and the increased risk of periodontal diseases. Vincent et al found that ND:YAG laser therapy is effective against EBV-positive periodontal diseases. A patient with refractory periodontitis and high EBV levels has significantly improved disease situation after receiving valacyclovir hydrochloride treatment. Most of the time, EBV is well controlled by human's immune system and is restricted in the status of latent infection. In people with immunodeficiency or some other specific situations, the latent infection of EBV will turn into lytic infection. The massive viral replication occurs, which leads to the death of host cells. The virion would be released into extracellular secreta, such as saliva. Grande et al. indicated that detections of EBV were more frequently in saliva and in subgingival plaque of HIV-positive patients than in HIV-negative periodontitis patients. However, no evidence has demonstrated whether lytic infection of EBV finally causes periodontitis, or periodontitis cause lytic infection happening. Therefore, more clinical application studies and further molecular mechanism researches are needed to determine the real relationships between EBV and periodontal diseases.The results of this meta-analysis also showed that EBV had severe associations with significant increased risks of different periodontal diseases, and these results may provide a new treatment strategy to substitute or assist traditional ways for periodontal disease treating, especially for late-stage disease. Some study groups tried to use physical or drug therapy to treat patients with EBV-positive periodontal diseases and found it was useful. Martelli et alThere are several limitations to current meta-analysis. First, the results of our meta-analysis were only applicable for CP and AgP because the associations between EBV and other types of periodontal diseases have not been reported. Second, it is possible to lead a language bias because all the inclusive studies were published in English and Chinese. Third, parts of subgroup analysis in this meta-analysis such as tissue, GCF samples, and site detecting included less studies, which may not provide a strong evidence to support the final results. Fourth, all the studies only referred to 3 ethnic groups—Europeans, Americans, and Asians, which may lead our meta-analysis limitative in application. Last, significant heterogeneity was detected between the studies included in quantitative synthesis. Through further subgroup analysis, we still cannot find the origin of heterogeneity. Hence, our results should be treated as exploratory and with caution.5Our meta-analysis based on 21 studies including 995 patients of periodontal diseases and 564 healthy people suggests that EBV is associated with increased risks of periodontitis including CP and AgP. In addition, this relationship exists in Asians, Europeans, and Americans. SgP and tissue are available for detecting EBV in patients of periodontitis. However, because of lack of enough evidence, detecting EBV in GCF sample still remains uncertain. At last, our results suggest that detecting EBV in samples from ≥5 (6) mm sites of periodontal pockets are more sensitive than ≤3-mm sites.


NQO1 C609T polymorphism and the risk of lung cancer, but an explicit consensus has not been reached. We conducted this updated meta-analysis to identify the function of NQO1 C609T polymorphism in lung cancer risk. All relevant literature was retrieved from the PubMed, EMBASE, CNKI, and WanFang databases before April 2017. A total of 37 studies (29 articles) with 8493 cases and 10,999 controls were included. Odds ratios (ORs) and 95% confidence intervals (CIs) were used to assess the strength of relations. We found that the NQO1 C609T polymorphism did not correlate with the risk of lung cancer in the overall analysis. In addition, no statistical significance was observed in the analysis stratified based on ethnicity, control source, quality score, or smoking status. A significant association was found in the subgroup of small cell lung cancer risk. Despite some limitations, this meta-analysis indicates that the NQO1 C609T polymorphism may not be associated with lung cancer risk. However, more epidemiological studies of larger samples and more ethnicities are needed to confirm these results.A variety of case-control studies have been performed to assess the correlation between  Worldwide, lung cancer is one of the most common cancers and the leading cause of cancer-related mortality among men, and the fourth leading cause of cancer morbidity and the second leading cause of cancer mortality among women , 2. In Cin vivo. By maintaining antioxidant forms of ubiquinone and reducing the endogenous quinones, NQO1 protects cells from oxidative damage [NQO1 gene, consisting of 6 exons and 5 introns, is located on chromosome 16q22. Numerous single nucleotide polymorphisms (SNPs) of the NQO1 gene have been identified, but the most frequently studied SNP is NQO1 C609T [NQO1 C609T  polymorphism is a C-to-T allele base-pair mutation in exon 6 (position 609) of the NQO1 gene. Such a gene mutation changes the encoded protein from a proline to a serine at position 187. This Pro187Ser mutated protein shows a reduced quinine reductase activity when compared with the wild-type protein [NAD(P)H: quinone oxidoreductase (NQO1), formerly called DT-diaphorase, is a flavoenzyme associated with carcinogen metabolism . As a twe damage . Researce damage . The humO1 C609T –12. NQO1 protein .NQO1 C609T polymorphism and lung cancer risk, but conflicting conclusions remain. In addition, several meta-analyses have been performed to reach a consensus but failed. To clarify the current uncertainty, we performed this comprehensive updated meta-analysis by incorporating all available literature.Epidemiological studies have assessed the correlation between the In the initial search stage, 81 potentially relevant published records were obtained. Through literature screening and abstracts reading, 38 articles were selected. We excluded 13 of those articles for the following reasons: 3 articles were case only research –16; 1 arNQO1 C609T polymorphism and lung cancer risk was observed in any of the genetic models .The calculated results of the meta-analysis are shown in Table Further subgroup analysis by ethnicity, control source, quality score, and smoking status still did not yield a significant association, except for the heterozygous model and dominant model in the Hawaiian subgroup  and the allele model in the quality score ≤ 9 subgroup . In the subgroup analysis by lung cancer subtype, statistically significant increased risks were found among non-small cell lung cancer for T vs. C  and small cell lung cancer for TT vs. CC , TT vs. CT+CC , T vs. C  and CT+TT vs. CC .Q-test and the I-squared statistics to test between-study heterogeneity. In the pooled analysis, significant heterogeneity exists among all five genetic models . Thus, the random-effect model was used to generate wider CIs. Moreover, we adopted a sequential leave-one-out sensitivity analysis to assess the influence of a single study on the combined ORs. After omitting each study, no substantial changes in ORs were observed, which suggested the robust and reliability of this meta-analysis .No evidence of obvious asymmetry in Begg's funnel plots was identified by visual observation Figure . In addiThe TSA showed that the cumulative Z-curve failed to cross the trial monitoring boundary before reaching the required information size, suggesting that more trials are needed to further verify the conclusions Figure .NQO1 C609T polymorphism in lung cancer susceptibility. The obtained results suggest that no significant relation exists. To date, this meta-analysis includes the largest samples used in the investigation of the function of NQO1 C609T in lung cancer susceptibility.In this meta-analysis, we investigated the role of NQO1 gene C609T polymorphism in lung cancer risk. In 1999, Chen et al. [NQO1 C609T polymorphism correlates with decreased lung cancer risk in Japanese. In another study conducted in England that consists of 82 lung cancer patients and 145 controls, Lewis et al. [NQO1 C609T and lung cancer risk. However, Masroor et al. [NQO1 609TT genotype could increase the risk of lung cancer in an Indian population of 100 lung cancer cases and 100 healthy controls. Several meta-analyses have been performed to obtain a clear correlation between NQO1 C609T and lung cancer risk. In a pilot meta-analysis conducted by Kiyohara et al. [NQO1 C609T polymorphism and lung cancer risk overall in African-Americans, East Asians, or Hispanics. However, the results suggested that NQO1 C609T polymorphism might correlate with lung cancer risk in Caucasians.Numerous studies have investigated the activity of the n et al.  claimed s et al.  failed tr et al.  reporteda et al. , only 10a et al.  failed tNQO1 C609T polymorphism and lung cancer risk. However, we failed to observe any significant relation between NQO1 C609T and lung cancer risk in the pooled analysis in any of the five genetic models. Subgroup analysis by ethnicity suggested that the T allele might be a protective factor in Hawaiians . In this Hawaiian subgroup, only one study, with 83 cases and 102 controls was included, which limits the strength of this result. Thus, more case-control studies in Hawaiians are needed to elucidate the role of C609T polymorphism in lung cancer risk. In another subgroup stratified by quality score, we did not detect any correlation between NQO1 C609T and lung cancer risk, yet the T allele acts as a risk factor for lung cancer in low-quality studies under the allele model . The insufficient statistical power of the relatively low-quality studies should be also considered. It is possible that such insufficient statistical power might result in false positive results. When we stratified the analysis on the basis of lung cancer subtype, an increased risk of the NQO1 C609T polymorphism on small cell lung cancer was observed. Previous pooled analyses have demonstrated that small cell lung cancer was more frequently observed in patients exposed to tobacco smoke [NQO1 C609T polymorphism might result in failure of NQO1 protein to detoxify highly toxic quinones, and thus have some consequences for smoking-related small cell lung cancer risk.This updated and comprehensive meta-analysis was performed to better elucidate the correlation between co smoke . The NQOIn this meta-analysis, we adopted many measurements to be certain of the credibility of our conclusion. First, we incorporated all eligible studies not only written in English but also written in Chinese to expand the included sample numbers. Then the sensitivity analysis and the publication bias were assessed according to the Cochrane protocol. The sensitivity analysis indicated that the results were robust, and Funnel plots suggested that no obvious publication bias was observed. The TSA was also used to test the conclusion reliability of this meta-analysis. The Z-curve failed to cross the trial monitoring boundary before reaching the required information size. Thus, more studies are needed to confirm or refute this finding. However, the limitations must be pointed out when the results of this meta-analysis are interpreted. First, we found that most of the comparisons show significant between-study heterogeneity, which might reduce the validity of the conclusion. Second, the strength of correlation was obtained by use of unadjusted estimates. Adjustment analysis was absent because of the lack of original data, such as environmental factors, age, drinking status, and gene-environment interactions, which restrains further analysis for risk factors. Nearly all the included case-control studies were conducted among Asians and Caucasians. Other ethnicities, such as Africans, were not studied. Additional studies are needed to confirm such conclusions from other ethnicities, especially Africans. Publication bias and language bias still exist, because only published studies and only the studies written in English or Chinese are included.NQO1 C609T polymorphism is not correlated with lung cancer risk, from the perspective of the former case-control studies. Further high-quality investigations with more detailed environmental exposure information and larger sample sizes are warranted to confirm our findings.The current meta-analysis provides evidence that NQO1 or quinone reductase or quinone oxidoreductase or DT-diaphorase or DTD,” and “lung cancer or lung neoplasm or lung carcinoma.” To enlarge the included studies, we also searched China National Knowledge Infrastructure (CNKI) and Wanfang databases using the above phrases in Chinese. In addition, eligible studies in the references of retrieved articles were also screened. The literature search was performed before April 2017 without any language publication restrictions. If an article contained two or more ethnic subpopulations, they were treated as separate studies. Only the largest study was included if two or more articles contained overlapping data. The designation and writing of this meta-analysis were under the guidelines of Preferred Reporting Items for Systematic Reviews and Meta-analyses.We performed a comprehensive literature search of PubMed and EMBASE by adopting the combination of the following phrases: “polymorphism or single nucleotide polymorphism or SNP or variant” and “NQO1 C609T polymorphism with lung cancer risk, (3) contained enough genotype distribution information to calculate ORs and 95% CIs, and (4) NQO1 C609T genotype frequency in control subjects were in agreement of the Hardy-Weinberg equilibrium (HWE).All the included articles in this meta-analysis met the following criteria: (1) contained unrelated case-control studies, (2) evaluated the correlation of Authors Jiawen Huang and Huiran Lin screened the articles and extracted data from all eligible studies, respectively. The data include first author's surname, publication year, country, ethnicity, the source of control subjects, genotyping methods, quality score, and numbers of cases and controls with CC, CT, and TT genotypes. Conflicting data were resolved by discussion after consensus was reachedThe quality of each study was assessed by use of the quality assessment criteria –32. The NQO1 C609T polymorphism and lung cancer risk. The strength of such correlation was assessed by calculating ORs with the corresponding 95% CIs. Stratification analyses were also performed by ethnicity, and source of control subjects, quality score, HWE in control subjects, smoking status, and lung cancer subtype. Between-study heterogeneity was analyzed by chi squared-based Q-test. When the studies were found to be homogeneous (with P > 0.10 for the Q-test), the fixed-effects model (the Mantel-Haenszel method) was used to estimate the pooled OR. Otherwise, the random-effects model (the DerSimonian and Laird method) was adopted. Sensitivity analysis was done by individually removing studies one by one and reanalyzing the pooled risk estimates. Begg's funnel plot and Egger's linear regression were used to estimate potential publication bias, with that asymmetric plot and a P-value < 0.05, indicating the presence of publication bias. HWE in the control subjects was tested by use of goodness-of-fit chi-squared test. A value of P < 0.05 was considered as departure from HWE. All statistical analyses were completed by STATA Version 11.0 software . All the statistics were two-sided, with significant findings set at a P-value of less than 0.05.All five genetic models, homozygous model (TT vs. CC), heterozygous model (CT vs. CC), recessive model (TT vs. CT + CC), dominant model (CT + TT vs. CC), and allele comparison (T vs. C) were adopted to investigate the correlation between To avoid the random errors caused by repeated significance testing and dispersed data, we performed the trial sequential analysis (TSA). First, the required information size was evaluated by considering an overall type I error (a) of 5% and type II error (b) of 20%. Then TSA monitoring boundaries were constructed on the basis of required information size as well as risk for type I and type II errors. If the cumulative Z-curve (blue line) crosses a TSA monitoring boundary (red lines) before reaching the required information size, the robustness of evidence might be confirmed and no further trials are necessary. Otherwise, to get a robust conclusion, more trials are needed.