Childhood maltreatment is associated with risky sexual behaviors (RSBs). Previous systematic reviews and meta-analysis focused only on the relationship between childhood sexual abuse and RSBs, thus the association between childhood maltreatment and RSBs has yet to be systematically and quantitatively reviewed. We aimed to provide a systematic meta-analysis exploring the effect of childhood maltreatment and its subtypes on subsequent RSBs in adolescence and adulthood. PubMed, Google Scholar, EMBASE, Medline were searched for qualified articles up to April 2019. We calculated the pooled risk estimates using either the random-effect model or fixed-effect model. The potential heterogeneity moderators were identified by subgroup and sensitivity analysis. Overall, childhood maltreatment was significantly associated with an early sexual debut (odds ratio (OR) = 2.22; 95% confidence interval (CI): 1.64–3.00), multiple sexual partners , transactional sex  and unprotected sex . Additionally, different types of childhood maltreatment were also significantly associated with higher risk of RSBs. Relevant heterogeneity moderators have been identified by subgroup analysis. Sensitivity analysis yielded consistent results. Childhood maltreatment is significantly associated with risky sexual behaviors. The current meta-analysis indicates it is vital to protect children from any types of maltreatment and provide health education and support for maltreated individuals. Childhood maltreatment (CM) including neglect and emotional, sexual, and physical abuse has been recognized as a global public-health and social-welfare problem ,2,3,4. ILikewise, risky sexual behaviors (RSBs) in adolescence and adulthood represent a serious public health concern that may lead to poor reproductive health outcomes such as sexually transmitted diseases (STDs), human immunodeficiency virus (HIV), unintended pregnancy and pelvic inflammatory diseases ,12. ChilFollowing these results, when exploring the effects of childhood maltreatment on risky sexual behaviors, different subtypes need to be considered. Previous systematic reviews and meta-analysis ,31,32,33We conducted the present meta-analysis strictly following the proposed PRISMA  statement. A systematic search of relevant articles in PubMed, Google Scholar, EMBASE, Medline was undertaken by two researchers. The following search terms were applied: “adolescent risky sexual behaviors”, “adult risky sexual behaviors”, “adolescent reproductive health outcomes”, “adult reproductive health outcomes”, “childhood maltreatment”, “child abuse”, “child adversity”, “child trauma”, “child neglect”, “child physical abuse”, “child emotional abuse”. We limited our searches to studies that were published between 1990 and 2018 and written in English, this time frame was chosen because the awareness of childhood maltreatment and its potential consequences was heightened globally during the time. We also manually searched review articles and reference lists of relevant studies. The exposure of interest was childhood maltreatment, which was further categorized into childhood sexual abuse (CSA), childhood physical abuse (CPA), childhood emotional abuse (CEA) and childhood neglect (CN). The outcomes of interest were risky sexual behaviors, including at least one of the following outcomes: early sexual debut, multiple sexual partners, transactional sex or unprotected sex. In our study, we focused not only on the association between total childhood maltreatment and any form of RSBs, but the association between subtypes of childhood maltreatment and any form of RSBs.We identified studies satisfying the following criteria: (1) population-based observational quantitative studies; (2) studies reported subtype(s) of childhood maltreatment such as childhood sexual, physical or emotional abuse, and/or neglect; (3) studies reported the associations between a history of childhood maltreatment and any form of risky sexual behavior; (4) studies included participants of both genders and reported the outcomes of the total sample; and (5) studies reported the relative risks (RRs) and odd ratios (ORs), with corresponding 95% confidence intervals (CIs) .Two independent researchers  extracted data and assessed study quality. The two researchers agreed upon any discrepancies through discussion and if necessary referred the issue to a third researcher. Using a study-designed standardized form, we extracted the available information as follows: first author, year of publication, study design, study country, total sample, subtype of childhood maltreatment and quality score. We used a ten-point Newcastle–Ottawa Scale (NOS) to assess the quality of included studies in the present review. Each study was judged on three broad perspectives: the selection of the study groups; the comparability of the groups; and the ascertainment of either the exposure or outcome of interest. When the study wins seven or more stars, it is considered of higher methodologic quality.Q statistics at the P < 0.10 level of significance. The I2 statistic, which is a quantitative measure of inconsistency across studies, was also calculated (significance level at I2 > 50%).We abstracted one effect size per outcome from each primary study. Odds ratios (ORs) with corresponding 95% confidence intervals (CIs) were used given the majority of the studies were cross-sectional. When evidence of heterogeneity existed across studies, we used random-effects models to calculate the pooled OR and the corresponding 95% CI, whereas fixed-effects models were used. When multiple types of childhood maltreatment were reported in the same study, the effect values of the relationship between different types of childhood maltreatment and the risk of having any form of risky sexual behaviors were combined with statistical software. Homogeneity of effect size across studies was tested by using the P < 0.10). All statistical analyses were performed using Stata version 13.0 and Review Manager version 5.3.A series of sub-group analyses according to geographic region, assessment method of CM, study design, study sample, study quality and year of publication were conducted to assess the potential effect modification of these variables on outcomes. We also conducted a sensitivity analysis by removing individual studies one at a time and recalculated a pooled OR to examine whether any one study overtly influenced the pooled effect size. We examined the possibility of publication bias, using Egger’s test and Begg’s rank correlation test   . However, significant heterogeneity was found ( < 0.01) . With re = 0.29) .I2 = 91%, P < 0.01) . Yet, substantial heterogeneity was found ( < 0.01) . With reectively .I2 = 84%; P < 0.01) , but not for CPA  . However, substantial heterogeneity was found ( < 0.01) . With re = 0.26) .I2 = 88%; P < 0.01) , but not in CPA  . Nevertheless, substantial heterogeneity was found ( < 0.01) . With re = 0.20) .I2 = 54.4%) was identified as the relevant heterogeneity moderator. However, the risk of having an early sexual debut was not statistically different for study design (P= 0.14). For risk estimates between childhood maltreatment and having multiple sexual partners, after subgroup analysis, study design (TSD: I2 = 90.6%), assessment method of CM (TSD: I2 = 67.1%) and geographic region (TSD: I2 = 2.8%) were identified as the first three most relevant heterogeneity moderators. The risk of having multiple sexual partners was statistically different for study design (P < 0.01). For risk estimates between childhood maltreatment and having transactional sex, after subgroup analysis, geographic region (TSD: I2 = 85.5%), study design (TSD: I2 = 1.1%) and quality score (TSD: I2 = 1.1%) were identified as the first three most relevant heterogeneity moderators. The risk of having transactional sex was statistically different for geographic region (P < 0.01). For risk estimates between childhood maltreatment and having unprotected sex, after subgroup analysis, study design (TSD: I2 = 87.3%), study sample (TSD: I2 = 71.7%) and geographic region (TSD: I2 = 59.2%) were identified as the first three most relevant heterogeneity moderators. The risk of having unprotected sex was statistically different for study design (P < 0.01).Subgroup analyses for risk estimates between childhood maltreatment and risky sexual behaviors are summarized in Sensitivity analyses were conducted to assess the stability of the results. We removed individual studies one at a time and recalculated a pooled OR to examine whether any one study overtly influenced the pooled effect size. Sensitivity analyses showed exclusion of any single study at a time did not materially alter the overall pooled estimates. For risk estimates between childhood maltreatment and having an early sexual debut, the pooled OR ranged from 1.91 (1.54–2.37) to 2.28 (1.55–3.34); for risk estimates between childhood maltreatment and having multiple sexual partners, the pooled OR ranged from 2.02 (1.66–2.46) to 2.33 (1.89–2.88); for risk estimates between childhood maltreatment and having transactional sex, the pooled OR ranged from 2.46 (1.80–3.34) to 3.46 (2.21–5.43); for risk estimates between childhood maltreatment and having unprotected sex, the pooled OR ranged from 1.44 (95% CI: 1.13–1.85) to 1.70 (95% CI: 1.27–2.29).P = 0.024) and Egger’s test (P = 0.020) indicated evidence of publication bias for risk estimates between CM and having an early sexual debut. Evidence of publication bias was observed by Egger’s test (P = 0.008), but not observed by Begg’s rank correlation test (P = 0.075) for risk estimates between CM and having multiple sexual partners. No indication of publication bias was observed by Begg’s rank correlation test  or Egger’s test  for risk estimates between CM and having transactional sex and having unprotected sex.Both the Begg’s rank correlation test  and having unprotected sex . Third, the association between childhood maltreatment and higher risk of RSBs still existed after subgroup and sensitivity analyses, indicating our results were stable and credible.Our meta-analysis yielded the following main findings. First, overall, childhood maltreatment was significantly associated with an increased risk of having an early sexual debut (OR = 2.22), multiple sexual partners (OR = 2.22), transactional sex (OR = 3.05) and unprotected sex (OR = 1.59). Second, different types of childhood maltreatment were also significantly associated with an increased risk of RSBs. For example, CSA (OR = 3.59), CEA (OR = 1.72) or CN(OR = 1.58) was significantly associated with risk of having an early sexual debut; CSA (OR = 2.63), CPA (OR = 1.54), CEA (OR = 1.74) and CN (OR = 1.44) also increased the risk of having multiple sexual partners; individuals experiencing CSA and CPA had a significantly higher risk of having transactional sex  gray matter and the alterations may possibly enhance sexual decision making and risky sexual behaviors ,57. MoreThe present meta-analysis has some limitations. First, the substantial heterogeneity among studies for the association between childhood maltreatment and RSBs were observed. However, this heterogeneity is not surprising given the differences in sociodemographic status and methodology. In our study, we conducted subgroup analyses to explain some extent of this heterogeneity and have identified the main heterogeneity moderators, including study design, assessment method of CM, geographic region, study sample and quality score. Although heterogeneity was still observed after subgroup or sensitivity analyses, the association between childhood maltreatment and RSBs still existed and the results were stable. However, many primary studies in our meta-analysis did not provide information on other variables which could explain some of the variance, such as duration of maltreatment, severity of maltreatment, the age of onset, perpetrator and comorbid disorders. Second, the majority of studies enrolled in the analyses were cross-sectional studies and retrospective cohort studies with the possibility of recall bias in self-reported childhood maltreatment and RSBs, thereby under- or over-estimating the strength of the association. However, evidence indicated the recall of childhood maltreatment remained fairly accurate over a long time. Third, discrepancies also existed in the definition of both CM  and RSBs, and the measurement of the constructs, frequency, and time frames of the outcomes, therefore our findings should be interpreted with caution. Fourth, many primary studies focused on school and community samples and few studies included street-involved youth and homeless individuals, thus our study may not be generalizable to other adolescent and adult populations. Finally, our study only included studies published in English, additional research in other populations is warranted to generalize the findings.Our study has important implications both for future research and prevention. To reduce the consequences of CM, we should be aware that not only child sexual abuse leads to RSBs, all forms of CM and other specific forms of CM  also contribute to RSBs. It is crucial to carry out research to clarify the unique effects of different forms of CM on RSBs and provide the evidence to take intervention measures in the future. However, out of 19 studies, only one of the included studies is from Europe, which indicates more sound research is supposed to be conducted in this valuable field. With regard to prevention, our findings highlight the need for effective programs and policies to screen for childhood maltreatment and provide more medical and reproductive health services for people who have experienced child maltreatment.In conclusion, the findings of our meta-analysis indicate that childhood maltreatment and its subtypes increase the risk of RSBs, although the potential bias and evidence of heterogeneity should be carefully considered. The current meta-analysis indicates that it is necessary to screen for maltreated history in childhood and provide health education and support for maltreated individuals, which have crucial implications for clinical practice and interventions.


HNF1B and cancer susceptibility, we conducted a comprehensive research synopsis and meta-analysis to evaluate associations between HNF1B variants and prostate and endometrial cancers.To provide a synopsis of the current understanding of the association between variants of OR = 1.247, p = 2.21 × 10− 77) and endometrial cancer ; the AA, AG genotypes also showed strong significant associations with the risk of prostate cancer . There was a strong significant association between rs7501939 G and the risk of prostate cancer . Strong significant association was found between rs11649743 G , rs3760511 C  and the prostate cancer risk;the GG, AG genotypes of rs11649743 also showed strong significant associations with the risk of prostate cancer . All the cumulative epidemiological evidence of associations was graded as strong.Eighteen studies totaling 34,937 patients and 55,969 controls were eligible for this meta-analysis. Four variants showed a significant association with the risk of individual cancer. Strong significant associations were found between rs4430796 A and the risk of both prostate cancer (HNF1B are associated with risk of prostate and endometrial cancer.Our study summarizes the evidence and helps to reveal that common variants of  Human cancers result in Considerable morbidity and mortality. Family history, ethnicity, lifestyle and region are potential risk factors for cancer development –4. HowevHNF1B, formerly known as TCF2) locus (on chromosome 17q12) and cancer risk [HNF1B deletion exhibit renal disease, elevated liver enzymes, and diabetes [HNF1B confers a poor overall survival in patients with multiple myeloma [Owing to advances in sequencing technologies and genome-wide association studies (GWAS), a large number of genetic variants correlated with various cancers have been identified , 9. Multcer risk –13. HNF1cer risk  and duricer risk . Patientdiabetes . HNF1B idiabetes , 17,ovardiabetes –20,endomdiabetes , 22 and  myeloma .HNF1B and the direction of their effects differ between cancer types. SNPs rs4430796 and rs7501939, are both associated with the prostate cancer risk across many ethnic groups [However, fine-mapping studies have revealed a complex genetic architecture of the HNF1B locus, demonstrating that variants of c groups . The samc groups . Yet, thc groups .HNF1B variants and cancer phenotypes, and performed a comprehensive meta-analysis, involving a total of 34,937 patients and 55,969 controls, to derive more precise estimates of the associations between HNF1B variants and susceptibility to prostate and endometrial cancers.Here, we collected data related to the associations between HNF1B variants and cancers published before July 2017. The search strategy was based on a combination of the terms (Hepatocyte nuclear factor-1 beta or HNF1B) and (cancers or tumors). The references of all computer-identified publications were searched for additional studies, and the PubMed option “Related Articles” was also used to search for potentially relevant papers. Searches were performed by two independent reviewers (Yu Tong and Yibin Wang). The language of the publications did not influence article selections.The US National Library of Medicine’s PubMed, Embase, OMIM, ISI Web of Science, and Chinese National Knowledge Infrastructure (CNKI) databases were searched in a systematic manner to retrieve all genetic association studies of HNF1B variants, and (3) the numbers of subjects possessing each allele and genotype in the cancer and control groups were available. No restrictions were set for the source of controls . Studies were excluded when: (i) they lacked sufficient information; (ii) they were published as letters to editors or conference abstracts; (iii) they were studies about cancer mortality.Studies were included if they met the following criteria. (1) the study reported original data from case-control or cohort studies, (2) the study reported alleles and genotypes for Data were extracted independently by two investigators (Yu Tong and Yibin Wang), who used recommended guidelines for reporting on meta-analyses of observational studies. The following data were extracted from the eligible studies: authors, journal title, year of publication, country of origin, selection and characteristics of cases and controls, demographic data, ethnicity of the study population, numbers of eligible and genotyped cases and controls, and genotype distributions in cases, controls, and available subgroups. Furthermore, we examined whether genotype frequencies in control groups conformed to the Hardy-Weinberg equilibrium (HWE) was determined. Any disagreement was adjudicated by a third author (Yi Qu).Minelli et al. We performed Cochran’s Q test and calculated І2 statistic to evaluate heterogeneity between studies. Harbord’s test was performed to evaluate publication bias. Potential small-study bias was evaluated by Egger’s test [metan, metabias, metacum, and metareg commands.The odds ratio was used as the metric of choice for each study. To detect overall genetic associations, allele frequencies were computed for studies reporting allele and genotype data. Pooled odds ratios were computed by the fixed effects model and the random effects model based on heterogeneity estimates. Once an overall gene effect was confirmed, the genetic effects and mode of inheritance were estimated using the genetic model-free approach suggested by r’s test . SensitiI2 < 25%, B for I2 between 25 and 50%, and C for I2 > 50%), and protection from bias . Cumulative epidemiological evidence for significant associations was thought to be strong if all three grades were A, moderate if all three grades were A or B, and weak if any grade was C.Venice criteria  were appTo determine whether a significant association could be excluded as a false positive finding, FPRP  was calculated using the method described by Wacholder et al. . FPRP < Our initial database search identified 113 potentially relevant studies. Based on a review of titles and abstracts, 55 articles were retained. The full text of these 55 articles was reviewed in detail, and 18 studies containing 36 datasets were eligible for inclusion in the meta-analysis. The specific process for identifying eligible studies and inclusion and exclusion criteria are summarized in Fig. Characteristics of the included articles are presented in Allelic associations: HNF1B rs4430796 and prostate cancer  and Caucasians  demonstrated a pattern similar to that of the full population. However, this effect was weak in the Africans . No publication bias was found in the eligible studies (Harbord’s test p = 0.253).All 15 publications were included in the evaluation of the association between the HNF1B rs7501939 and prostate cancer . No publication bias was found in the eligible studies (Harbord’s test p = 0.864).Six publications were included in the evaluation of the association between the HNF1B rs11649743 and prostate cancer . Strong significant association was observed with the prostate cancer risk .Two publications included data regarding the association between the HNF1B rs3760511 and prostate cancer. There was a strong significant association between rs3760511 and the risk of prostate cancer, and moderate heterogeneity was found among the eligible studies . No publication bias was found in the eligible studies (Harbord’s test p = 0.778).Three publications were included in the evaluation of the association between the HNF1B rs4430796 A > G and endometrial cancer . Lack of significant association was found in Africans , the Asians , and Latino and Hawaiian . No publication bias was found in the eligible studies (Harbord’s test p = 0.950).Two publications were included in the evaluation of the association between the HNF1B rs7501939 and endometrial cancer . The same pattern was observed in Caucasians , Africans , Asians and Latino and Hawaiian  (Data not shown).Two publications were included in the analysis of the association between the HNF1B rs4430796 among case and control groups is presented in Table OR1) and AG versus GG (OR2) were calculated for each study. A multivariate meta-analysis was conducted to estimate the pooled risk. There was a significantly increased risk of prostate cancer among individuals with the homozygous AA genotype  and heterozygous AG genotype .The pooled estimates were similar to those obtained after removal of the study with HW disequilibrium [OR1 = 1.524  and random effect OR2 = 1.198 .Of the 15 publications, only seven reported genotype information. The genotype distribution of the HNF1B rs11649743 among case and control groups is presented in Table OR1) and GA versus AA (OR2) were calculated for each study. Multivariate meta-analysis was conducted to estimate the pooled risk. There was a significantly increased risk of prostate cancer among individuals with the homozygous GG genotype  and heterozygous AG genotype . No between-study heterogeneity was found for the homozygous GG genotype  or for the heterozygous GA genotype .Only one publication reported genotype information for rs11649743. However, this publication included relevant data for different populations and regions. The genotype distribution for the Venice criteria were applied to evaluate these significant associations. Details of protection from bias for genetic variants significantly associated with prostate and endometrial cancer risk in meta-analyses are shown in Table P value was calculated. All associations with cancer risk had a FPRP value < 0.001. Thus, all the cumulative epidemiological evidence of associations was graded as strong.To evaluate the probability of true association with cancer risk for the nominally significant variants, FPRHNF1B variants and susceptibility to prostate and endometrial cancers. Our primary analysis revealed that, rs4430796 A, showed strong significant associations with risk of both prostate cancer  and endometrial cancer ; the AA, AG genotypes also showed strong significant associations with risk of prostate cancer . Sensitivity analyses in Caucasians demonstrated patterns similar to that of the full population. However, lack of significant association was found in Africans, which is likely due to the considerably smaller sample size. There was a strong significant association between rs7501939 A and the risk of prostate cancer ; however, lack of significant association with endometrial cancer risk was observed , and the GG, AG genotypes also showed strong significant associations with the risk of prostate cancer . Strong significant association was also found between rs3760511 C and the risk of prostate cancer . Using the Venice criteria and false-positive report probability tests, we graded all the cumulative evidence of significant associations with prostate and endometrial cancers risk as strong.To our knowledge, this is the first general overview of the association between Our findings were based on several gene-association studies, including several thousand participants, and were robust in terms of study design and sensitivity analyses. We found no evidence of publication bias or small study bias based on funnel plots. Between-study heterogeneity was found in allelic association studies (G versus A) of rs7501939, and in allelic (G versus A) of rs11649743 for prostate cancer. When HWE was examined, one study showed deviation. Our results were robust to the removal of this study.HNF1B encodes three isoforms: isoforms ; isoform A and B act as transcriptional activators and isoform C acts as a transcriptional repressor [HNF1B might modulate the risk of cancer [HNF1B were associated with the risks of both endometrial cancer in women of European background [HNF1B rs757210 is specific to serous epithelial ovarian cancer by pooling data from GWAS and follow-up genotyping; the analysis included 43 studies from the Ovarian Cancer Association Consortium [HNF1B rs7501939 was a susceptibility locus for testicular germ cell tumor [HNF1B variants predispose individuals to clear cell ovarian, endometrial, lung and prostate cancers, et al.epressor . HNF1B if cancer . Howeverckground  and prosckground  . Severackground , 46, 47.ckground  . In 201nsortium . At the ll tumor . Taken tThere are several limitations of the study. First, it is likely that some publications were overlooked although we conducted an exhaustive literature search, some relevant published studies with null results were not identified. Second, due to insufficient data, we were unable to evaluate publication bias for associations between several variants in 8q24 region and prostate and endometrial cancer. Third, a unified analysis standard across studies could not be defined for lack of raw data from the original publications. Therefore, future studies with larger sample size are warranted to confirm these associations.HNF1B variants to cancer biology, we attempted to estimate the strength of the genetic associations between these variants and prostate and endometrial cancers. This Human Genome Epidemiology (HuGE) systematic review presents strong evidence for an association between HNF1B variants and prostate and endometrial cancers, both overall and in Caucasians, Asians, Africans, and Indians, suggesting a multiplicative genetic model for variants of HNF1B among different ethnic populations. Our study results also suggest that HNF1B plays an important role in prostate and endometrial cancers, and these variations may serve as efficient and economical biomarkers for the diagnosis of prostate and endometrial cancers.Given the relevance of


IGF1 gene, rs2195239 and rs2162679 attracted many attentions. So we perform a meta-analysis to determine potential associations between IGF1 rs2195239 and rs2162679 polymorphisms and cancer risk.Many epidemiological studies have suggested that insulin-like growth factor1 (IGF1) gene single-nucleotide polymorphisms (SNPs) may be associated with cancer risk. Among several commonly studied polymorphisms in We retrieved relevant articles from the PubMed, Embase, and Web of Science databases up to April 30, 2018. Ultimately, thirteen studies were included in the present meta-analysis, which involved 12,515 cases and 19,651 controls. The odd ratios (ORs) and their 95% confidence intervals (CIs) were pooled to estimate the strength of the associations.rs2195239 reduces the overall cancer risk in homozygote model, as well as reducing cancer risk in Asian populations in allele, homozygote, and recessive models. No significant relationship was found between rs2195239 and breast or pancreatic cancer risk. rs2162679 reduces the overall cancer risk in allele, homozygote, dominant, and recessive models, as well as reducing cancer risk in Asian populations in allele, homozygote, and recessive models.IGF1 rs2195239 and rs2162679 were associated with overall cancer risk based on present studies.The online version of this article (10.1186/s12881-019-0749-3) contains supplementary material, which is available to authorized users. Insulin-like growth factor1 (IGF1) plays an important role in promoting cell proliferation and inhibiting apoptosis . EpidemiIGF1 is located on 12q22–24.1, having no strong linkage disequilibrium with nearby genes [IGF1 SNPs have been reported to be associated with elevated IGF1 levels in the circulation [by genes . Studiesby genes –10. Singculation –13.IGF1 gene rs2195239 and rs2162679 polymorphisms and the risks of various cancers [IGF1 rs2195239 and rs2162679 polymorphisms and cancer risk.Recently, many studies have described the relationship between the  cancers –23. HoweRelevant articles published prior to April 30, 2018 were identified by searching the PubMed, Embase, and Web of Science databases. The key terms used in the search were: “IGF1 or IGF-1 or insulin-like growth factor 1,” “variant or mutation or SNP or polymorphism,” and “cancer or tumor or neoplasm or carcinoma.” Furthermore, we manually checked the reference in the identified articles to identify additional available studies. Our search was restricted to articles written in the English language.IGF1 polymorphisms rs2195239 and rs2162679 and cancer risk, 2) be case-control or cohort studies, 3) contain sufficient data on genotype distribution. We excluded comments, editorials, reviews, meta-analyses, and studies lacking sufficient data.The included articles had to: 1) concern the relationship between the P-value for the Hardy-Weinberg equilibrium (HWE) of controls. The quality of the studies was assessed using a quality score form [Two researchers extracted information from all the included studies independently, as well as evaluating the quality of the studies. Controversies were resolved through negotiation. The following data were collected: first author’s name, publication year, type of cancer, ethnicity, method of genotyping, control source, genotype distributions of cases and controls, and the ore form  , the homozygote model , the heterozygote model , the dominant model , and the recessive model . Heterogeneity was estimated using a Q test and I2 [P > 0.1). Otherwise, the random-effects model were used [Statistical analyses were applied by using STATA software . The ORs and 95% CIs were calculated to evaluate the strength of the associations between the t and I2 . The fixere used . The HWEere used .IGF1 polymorphisms [Through searching the databases, a total of 4479 articles were initially obtained. After removing duplicates, 2086 articles were left. After screening the titles and abstracts, 133 articles were retrained for full-text review. Ultimately, ten articles were identified for meta-analysis –23 . In the analyses stratified by ethnicity, rs2195239 was shown to significantly reduce cancer risk in Asian populations . In the analyses stratified by cancer type, no significant relationship between rs2195239 and breast  risk was found. The scores for all of the studies regarding rs2195239 are no less than twelve.There were a total of 9842 cases and 14,105 controls included from eight studies regarding the rs2195239 polymorphism. In overall analysis, rs2195239 was shown to be significantly associated with reduced cancer risk . In the analyses stratified by ethnicity, we found that rs2162679 was shown to significantly reduce cancer risk in Asian populations . The results synthesized from these studies, which score no less than twelve, showed that rs2162679 reduces cancer risk in the allele and homozygote models  and dominant models  of the SNPs were shown to be higher than 20% among most of the populations , and there was no close LD between rs2162679 and rs2195239  and found that the conclusions remained the same  and rs2162679 (genomic position: Chromosome12: 102477481) and polymorphisms in the populations from the 1000 Genomes Project Phase 3. Table S4. Meta-analysis of the association between rs2195239 polymorphism and cancer risk, omitting the study of Patel or Birmann. Table S5. The OMIM numbers for important genes and pathogenic conditions in this study. Figure S1. Linkage disequilibrium analyses for IGF1 rs2195239 and rs2162679 polymorphisms in populations from the 1000 Genomes Project Phase 3. 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Previous studies regarding the relationship between acne and prostate cancer risk have reported inconsistent results. We performed the present meta-analysis of observational studies to summarize the evidence on this association.2 index were used to evaluate the heterogeneity across the studies.A comprehensive literature search up to March 2018 was performed in PubMed, Scopus, Web of Science, and Chinese National Knowledge Infrastructure (CNKI) databases. Summary odds ratios (ORs) with 95% confidence intervals (CIs) were estimated with a random effects model. The Q statistic and the I2 = 64.5%). In the subgroup analysis by study design, a significant association was observed in the cohort studies  but not in the case-control studies .Eight studies were ultimately included in this meta-analysis. In the overall analysis, no significant association was found between acne and prostate cancer risk . A significant heterogeneity was observed across studies (P = 0.006, IIn summary, this meta-analysis did not find an association between acne in adolescence and prostate cancer risk. However, because there was some heterogeneity in the overall analysis and a significant association was observed in the meta-analysis of the cohort studies, further well-designed large prospective studies are warranted to confirm our results. Prostate cancer is the second most commonly diagnosed cancer in men worldwide, with 1.1 million new cases estimated to have occurred in 2012 . TherefoPropionibacterium acnes (P. acnes), a skin bacterium closely associated with acne, is reported to be associated with prostatic inflammation and carcinogenesis [Recently, emerging studies have examined the association between acne and the risk of prostate cancer given that acne vulgaris is a proxy for androgen status . In addiogenesis –9. Severogenesis  indicateogenesis . By contogenesis  and Cremogenesis  failed tA comprehensive literature search was performed in PubMed, Scopus, Web of Science, and Chinese National Knowledge Infrastructure (CNKI) databases up to March 2018 with the following search keywords: (acne or propionibacterium) and (prostatic neoplasms or prostatic cancer or prostate neoplasms or prostate cancer). We first evaluated the potentially relevant articles by checking their titles and abstracts. Any studies possibly matching the eligibility criteria were further evaluated by reading the full texts. We also examined the cited references lists from retrieved articles and reviews to identify any additional relevant studies. This systematic review and meta-analysis was planned, performed, and reported in adherence to the reporting and publication requirements of the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) statement .Studies included in this meta-analysis met all of the following criteria: (i) the exposure of interest was acne; (ii) the outcome of interest was prostate cancer; (iii) the study had a cohort, nested case-control, or case-control design; and (iv) the study provided risk estimates with their corresponding 95% confidence intervals (CIs) or enough data to calculate them. There was no restriction on language or publication date.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp). The NOS is an eight-item instrument that primarily assesses the following sections of each study: source of study population, study comparability, follow-up, and outcome of interest. We assigned NOS scores of <7 and ≥7 for low- and high-quality studies, respectively.The method quality of each study was assessed by two independent reviewers (X.Z. and Y.L.) with the Newcastle-Ottawa Scale  with a predefined information collection form. Any discrepancies were resolved by group consensus. We recorded the following data for each study: first author’s surname, the country in which the study was performed, publication year, study design, sample size, methods of exposure assessment, fully adjusted risk estimates with their 95% CIs, and adjusted confounders in the data analysis.As prostate cancer is a rare disease, the relative risk (RR) and the hazard ratio (HR) were assumed to be approximately the same as the odds ratio (OR). The ORs and their 95% CIs were used as the study outcome to assess the strength of the relationship between acne and prostate cancer risk. Summary ORs with 95% CIs were estimated with a random effects model using the method of DerSimonian-Laird , which i2 score (95% CI) was further used to assess the degree of heterogeneity  [Heterogeneity across studies was evaluated using the Q statistic with a significance level set to P value < 0.10 . I2 scorgeneity) . Finallygeneity)  were useSensitivity analysis was performed by repeating the pooled analysis after omission of each study in turn. A cumulative meta-analysis was performed by sorting each study by publication year. All of the statistical analyses were performed with Stata 10.0 . A two-sided P value < 0.05 was considered statistically significant.P. acnes antibody were also excluded. Eight studies [The detailed process of our literature search and selection is shown in  studies , 18–21 w studies , 11, 20  studies , 21. TheThe multivariable-adjusted ORs for each study and combination of all studies for acne versus no acne groups are shown in 2 index were used to evaluate the heterogeneity across the studies. Significant heterogeneity was observed among studies in the overall analysis (2 = 64.5%). Using Galbraith plot analysis, the study by Ugge et al. [In this meta-analysis, the Q statistic and the Ie et al.  was idenA sensitivity analysis was performed to assess the influence of each study on the overall estimate. Meta-analysis estimates were computed by omitting one study at a time. The overall estimate (95% CIs) ranged from 1.03 (95% CI 0.89–1.20) to 1.12 (95% CI 0.96–1.31) after the omission of the study by Ugge et al.2018)  and the   and theA cumulative meta-analysis was performed on the included studies sorted by publication year. As shown in The present study summarized the current evidence of the relationship between acne in adolescence and prostate cancer risk using a meta-analysis of observational studies, including five case-control studies and three cohort studies. To the best of our knowledge, this is the first meta-analysis evaluating the association between acne and prostate cancer risk. The results indicated that acne was not significantly associated with the risk of prostate cancer .When we assessed heterogeneity across the studies with a meta-regression method, we found that study design was a potential source of heterogeneity in the overall meta-analysis. The summary OR (95% CI) for the cohort studies was 1.51 (1.19–1.93) without obvious heterogeneity across the studies. However, no significant association was observed in the case-control studies, and this meta-analysis of 8 studies was outweighed by the case-control studies, which would tend to make the summary risk estimate skew toward the null. However, as cohort studies are generally superior to case-control studies, further large prospective cohort studies are warranted because of the positive findings in the subgroup of cohort studies.This study had some strengths. A total of 10,145 prostate cancer cases were included in this meta-analysis; the large sample size provided reliable summary risk estimates. We extracted data from the most fully adjusted models in each study, which might minimize potential confounding factors. Various subgroup analyses, heterogeneity analyses, and sensitivity analyses were performed to evaluate the robustness of our findings.P. acnes isolated from radical prostatectomy specimens have been reported to be positively related to the onset and extent of both acute and chronic prostate inflammation [P. acnes were more common in men with prostate carcinoma than in the controls, and the cells infected with P. acnes had increased proliferation in vivo [P. acnes infection may be a contributing factor to the development of prostate cancer [P. acnes. However, in this study, we failed to detect a potential link between a history of acne and prostate cancer risk.The exact pathogenic mechanism of prostate cancer is still largely unknown. It has been hypothesized that chronic inflammation may contribute to prostate carcinogenesis . P. acneammation . P. acne in vivo , 24, 25.e cancer , 27. In e cancer  and hypee cancer ), into p2 = 64.5%), which might distort the pooled risk estimates. This is a large meta-analysis, and there is a direct link between meta-analysis size and detected heterogeneity. Thus, obvious heterogeneity is the norm, and it is great if heterogeneity has been identified and can be incorporated in the model [This study had several limitations that should be acknowledged. First, substantial heterogeneity was detected across individual studies Click here for additional data file.S1 Fig(TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.


GST gene polymorphism and red meat intake on CRC occurrence in the Polish population were analyzed in this study. Methods: Genotyping was performed with the qPCR method. Results: A high frequency of meat consumption was associated with an over 2-fold increase in the risk of colorectal cancer odds ratio (OR) adjusted for sex and age = 2.4, 95% confidence interval (CI); 1.3–4.4). However, after analyzing the genetic profiles, in the absence of polymorphisms of all three analyzed genes, there was no association between a high frequency of meat consumption and the occurrence of CRC. In the case of GSTM1 gene polymorphism, the high frequency of meat consumption increased the risk of CRC by almost more than 4 times . For GSTP1 gene polymorphism, a 3-fold increase in CRC risk was observed with a high frequency of meat consumption . In the case of GSTT1 gene polymorphism, the increase in risk of CRC was not statistically significant . Conclusions: The frequency of red meat intake in non-smokers increases the risk of colon cancer in the case of GST gene polymorphisms.Background: It is postulated that both individual genotype and environmental factors such as diet may modify the risk of developing colorectal cancer (CRC). The influences of  Colorectal cancer (CRC) is the third most commonly diagnosed cancer and the second most deadly cancer worldwide. According to GLOBOCAN 2018 data, there were 1.8 million new cases of CRC and nearly 1 million deaths due to this disease in 2018 [A high percentage of colorectal cancer cases are found in developed countries with a high degree of industrialization, where the society is characterized by a so-called Western lifestyle. A diet rich in red processed meat seems to be an integral element of this lifestyle ,4. Thus,The mechanism underlying the potential carcinogenicity of red meat has not been entirely explained so far. Here, several potential mechanisms are postulated. Most attention is focused on heterocyclic amines (HAA) as well as polycyclic aromatic hydrocarbons (PAH), which arise during the thermal processing of red meat . These cIt is noteworthy that in processed meat with a high content of salts, nitrates, and nitrites, there are N-nitroso compounds with mutagenic and carcinogenic potential. In addition, these compounds are endogenously formed as a result of meat digestion . A high In a prospective study of the American population by the National Institutes of Health (NIH) including around 0.5 million people, it was confirmed that more frequent red meat consumption increases the risk of CRC; however, this relationship was stronger for rectal cancer than for colon cancer . A meta-Despite the fact that red meat has been recognized as a cancer-promoting factor, the results of studies do not always explicitly confirm the relationship between the consumption of red and processed meat with the risk of gastrointestinal cancers ,9. A pooGST), which are responsible for detoxification processes, including antioxidant activity, belong to the group of genes whose polymorphisms may be crucial for carcinogenesis. They represent a superfamily of genes that encode phase II metabolizing enzymes for a number of toxic substances. Their substrates are, among others, heterocyclic amines, which are potential carcinogens in red meat subjected to thermal treatment. In addition, the polymorphism of GST genes results in a decrease or a complete lack of enzymatic activity of coded transferases [The lack of unequivocal conclusions may result from different types and patterns of research or from the limitations of experiments analyzing the frequency of consumption (food frequency questionnaires (FFQs)) . Perhapssferases . TogetheGST gene polymorphism and the frequency of red meat consumption in the risk of developing colorectal cancer in a Polish population. Confirmation of possible dependencies between the high consumption of red meat in people with decreased detoxification activity resulting from GST gene polymorphism and the risk of developing CRC would allow better future adaptation of prophylactic recommendations to this group of patients.Based on this hypothesis, the aim of the study was to analyze the correlation between The recruitment of patients participating in the study took place at two clinical centers in Kielce, Poland, the Provincial Hospital in Kielce and the Holy Crosss Cancer Centre, in 2014–2017. Current and former smokers were excluded from the population. The same carcinogenic compounds are contained in tobacco smoke that arises from the heat treatment process and/or the passage of red meat through the digestive tract. Thus, smoking by patients disrupts the assessment of the impact of red meat intake in the context of exposure to carcinogenic substances such as HAA, PAH, ROS, and N-nitroso compounds. The study group included 197 patients with confirmed colorectal cancer based on a pathomorphological diagnosis of specimens collected during colonoscopy or surgery. The control group consisted of 104 patients without cancer, as confirmed by an endoscopic and/or histopathological examination. All patients signed written consent forms prior to participation in the study. The study was approved on 3 June 2013 by the local Bioethics Commission (No. 5/2013) on the basis of the submitted application with an exact description of the procedure.The frequency of red meat consumption was assessed with the FFQ. The questions concerned the usual frequency of red meat consumption, including the consumption of beef, pork, veal, mutton, and venison, categorized by the method of preparation. It is noteworthy that each question covered one kind of meat dish, without differentiation for particular type of red meat  to be as respondent friendly as possible. Questions 1–5 concerned the frequency of consumption of red meat subjected to thermal treatment , whereas questions 6–8 concerned the frequency of processed red meat consumption in the form of cold meat, canned food, and sausages. The questionnaire was completed independently by the patients after prior instructions from a nurse trained to participate in the project.GSTT1 (Assay ID Hs00010004_cn) and GSTM1 (Assay ID Hs02575461_cn) was conducted using the qPCR relative quantification method. The TERT gene constituted the internal control of reaction. Analysis of the GSTP1 gene SNP (rs1695) polymorphism was conducted using TaqMan qPCR endpoint genotyping (Assay ID C_3237198_20). In all cases, the Light Cycler 96 instrument and TaqMan primer/probe kit  were used.Peripheral blood leukocytes were used for genetic testing. DNA was isolated from frozen blood samples using the Genomic Micro AX Blood Gravity kit from AA Biotechnology . The purity and concentration of isolated DNA were evaluated spectrophotometrically at 260 nm and 280 nm  Waltham, MA, USA). Analysis of the U test due to non-normality which was assessed by the Shapiro–Wilk test. Crude and adjusted odds ratios (OR) with 95% confidence intervals (95% CI) were calculated in a logistic regression model. A two tailed p-value < 0.05 was considered statistically significant. All statistical analyses were performed using R  and Statistica , the distribution of polymorphisms in lication . Additiolication ,14. Our GST gene polymorphism in the risk of developing CRC. The results of this analysis are presented in GSTM1 gene polymorphism, a high frequency of meat consumption increased the risk of CRC by almost 4 times. Furthermore, for polymorphism of the GSTP1 gene, a 3-fold increase in CRC risk was observed with a high frequency of meat intake. In the case of the GSTT1 gene polymorphism, a high frequency of meat consumption increased the chance of CRC occurring in a statistically insignificant manner, which may have resulted from the small size of the group (52 people) (The study examined the correlation between the frequency of red meat consumption and the  people) .Cytosolic glutathione S-transferase genes seem to be highly polymorphic in humans. They occur in dozens up to hundreds of variants observed in populations around the world with varying frequency. Polymorphisms with a high or even 50% frequency of occurrence in Europe were analyzed in the study. The frequency of analyzed polymorphisms obtained for the Polish population confirmed the data acquired by other researchers from Europe ,15,16.The analyzed polymorphisms reduce the detoxification capacity of this group of enzymes in theory, but, in fact, there was no statistically significant relationship between the genetic profile of individuals and the risk of developing CRC. Therefore, the aim of the project was to investigate the interaction between genetic factors and the influence of the environment in the form of specific eating habits in modulating the risk of colon cancer. Red and processed meats seem to be elements of the diet with highly unfavorable effects on the intestinal mucosa of people with reduced detoxification potential. The International Agency for Research on Cancer (IARC), a member of the World Health Organization (WHO) within the European Code Against Cancer, recommends reducing the consumption of red and processed meat in the diet, especially due to the risk of CRC .According to Jarosz and coauthors, the consumption of this product group was the highest in Poland in the late 1970s and early 1990s, which could have influenced currently diagnosed cancer cases . It is pp = 0.005) between the patients and controls was presented in terms of the number of people declaring a high consumption of red meat . Based on the above data, it can be assumed that very frequent consumption of red meat may potentially increase the risk of developing colorectal cancer. The probable mechanism conducive to carcinogenesis may be the regular exposure of mucous membrane to toxins generated in the digestion process of this food product, such as HAA, nitrosamines, ROS, or increased oxidative stress resulting from a high content of heme iron [GST genes, it was found that such a risk increases only in people with a deficiency of glutathione S-transferases resulting from mutations of the genes encoding them.A statistically significant difference  and GSTP1 (p = 0.018) genes. The studied polymorphism of the GSTM1 gene is based on a deletion of both gene alleles (genotype null/null) and results in a lack of protein product, so it can be assumed that deficiency of this detoxification enzyme glutathione S-transferase µ1 in connection with increased exposure to carcinogenic substances of red meat is a risk factor for CRC development.Statistically significant differences in the frequency of red meat intake between the patients and controls were observed for people with polymorphism of GSTM1 gene polymorphism, a high frequency of meat consumption resulted in the highest risk of the three analyzed genes, almost a 4-fold increase in the risk of CRC. In fact, it is believed that glutathione transferase μ1 shows higher efficiency in the metabolism of xenobiotics than other GST enzymes and plays a key protective role against cytotoxic and genotoxic compounds in the gastrointestinal tract [According to our results , in the al tract .GSTP1 gene  resulted in the expression of a protein with reduced enzymatic activity. An amino acid change at position 105 of the polynucleotide chain from isoleucine to valine affects the geometry of the substrate binding area by GSTP1 transferase [The analyzed polymorphism of the nsferase . This mogene SNP 5 Ile/ValGST polymorphism did not show statistically significant differences between patients and controls, which means that high red meat intake in people with a normal enzyme apparatus of the glutathione S-transferase family does not affect the risk of colon cancer. It can be assumed that despite the regular consumption of red meat, exposure to carcinogens such as HAA, N-nitroso compounds, and ROS is minimized thanks to efficient detoxification systems with the participation of GST transferases. Analysis of the frequency of red meat intake in people without the There are some limitations in our study. The FFQ survey used to assess the frequency of consumption of red meat did not allow a quantitative assessment of the consumption of red meat by patients. Nevertheless, a meta-analysis of studies from Europe, North America, and Australia has shown that the regularity and high frequency of eating red meat is more important for the risk of developing CRC than the daily amount . There wGST gene polymorphism and high (over once per day) consumption of red and processed meat. Such features are exhibited by as many as 69.1% of controls without changes in the large bowel. It is therefore considered that this portion of the controls should have constant monitoring. They have a 3-fold higher than normal risk of developing colorectal cancer in the future. It is advisable to include those people in further prospective research on the subject, which will be the goal of project continuation.There are some clear conclusions from the study. First of all, the red meat intake in the Polish population is very high. The study showed an increased risk of CRC in people with a


Several studies have investigated the association between Toxoplasma gondii (T. gondii) infection and risk of Parkinson's disease (PD) with inconsistent results. Clarifying this relation might be useful for better understanding of the risk factors and the relevant mechanisms of PD, thus a meta-analysis was conducted to explore whether exposure to T. gondii is associated with an increased risk of PD. We conducted this meta-analysis according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement. A rigorous literature selection was performed by using the databases of PubMed, Embase, Web of Science, Cochrane Library, and ScienceDirect. Odds ratio (OR) and corresponding 95% confidential interval (CI) were pooled by using fixed-effects models. Sensitivity analysis, publication bias test, and methodological quality assessment of studies were also performed. P=0.314) and acute infection  were not associated with PD risk. Stable and robust estimates were confirmed by sensitivity analysis. No publication bias was found by visual inspection of the funnel plot, Begg's, and Egger's test. Seven studies involving 1086 subjects were included in this meta-analysis. Pooled data by using fixed-effects models suggested both latent infection (OR, 1.17; 95% CI, 0.86 to 1.58; This meta-analysis does not support any possible association between T. gondii infection and risk of PD. Researches are still warranted to further explore the underlying mechanisms of T. gondii in the pathogenesis of PD and their causal relationship. With an estimated prevalence of 315 per 100,000 worldwide, Parkinson's disease (PD) is the second most common neurodegenerative disorder next to Alzheimer's disease . For theToxoplasma gondii (T. gondii) is an intracellular protozoan parasite that causes a zoonotic disease known as toxoplasmosis. Over thirty percent of the world population was infected with T. gondii, and it is viewed as the most prevalent infection in humans . PrimariSo far, inconsistent conclusions of several epidemiological studies investigating the association between T. gondii infection and PD risk have been reported. Ramezani et al.  suggesteA systematic literature search was conducted in five English electronic databases from their inceptions to October 2018. The databases included PubMed, Embase, Web of Science, Cochrane Library, and ScienceDirect. Keywords for literature search were listed as follows: “Toxoplasma gondii”, “toxoplasmosis”, “Parkinson's disease”, and “Parkinsonism”. Synonyms and variations of keywords were applied to ensure a comprehensive search. Reference lists of relevant reviews were also screened for potentially eligible literature. The detail search strategy for PubMed database was available in Two reviewers independently screened the titles and abstracts of retrieved articles. For publications whose eligibility could not be determined by the above process, full-text reading was performed for further evaluation. Studies meeting the following criteria were considered eligible: (a) case-controlled or cohort studies investigating the relationship between T. gondii infection and risk of PD; (b) the exposure was T. gondii infection diagnosed by positive serum anti-T. gondii Ig G antibodies or IgM antibodies, which respectively represented latent and acute infection ; (c) theStudies were excluded if they were (a) reference papers, case reports, reviews, letters, and methodological or nonpopulation-based researches; (b) studies discussing nontargeted outcomes or exposures; (c) studies with incomplete data even if attempts were made to contact authors.Any disputes were settled by discussion or referral to a third reviewer.The related data of included studies were independently extracted by two investigators, with a standardized data extraction checklist. Extracted data of individual study included first author, publication year, region, study design, number of cases and controls, demographic features of participants , measurement of exposure, HR, RR, or OR with corresponding 95% CI, adjusted factors.In this review, Newcastle-Ottawa quality assessment scale (NOS)  was applDisagreements regarding data extraction and methodological quality appraisal were resolved by a third investigator. Q statistic, with a significant level of 0.1. We also evaluated statistical heterogeneity by using I2 statistic, whose cut-off values of 25%, 50%, and 75% were, respectively, viewed as low, medium, and high heterogeneity [The association between T. gondii infection and risk of PD was measured by OR. The HR and RR were considered to be equivalent to OR for the low incidence of PD in the population . Heterogogeneity . We syntogeneity . SensitiLiterature screening process and corresponding exclusion reasons were illustrated in Of the seven studies, three originated from Iran , 26, 29,I2=40.5%, P=0.108), results of pooled fixed-effects analysis indicated latent infection of T. gondii was not associated with risk of PD . After excluding low-quality studies, sensitivity analysis demonstrated the similar results , suggesting pooled results were stable and robust.Seven studies , 37, 38 I2=0.0%, P=0.490); a fixed-effects model was used to synthesize data. Pooled results suggested no association was observed between acute infection of T. gondii and PD risk .Three studies , 26, 28 Z=0.37, P=0.711) and Egger's linear regression test  nor acute infection  was associated with risk of PD.Our meta-analysis systematically reviewed all existing studies investigating the association between T. gondii infection and risk of PD, and seven case-controlled studies involving 1086 subjects were included in the pooled analysis. We found neither T. gondii latent infection (OR, 1.17; 95% CI, 0.86 to 1.58;To the best of our knowledge, this is the first meta-analysis investigating the association between T. gondii infection and PD risk. Eligible studies were selected by applying comprehensive and rigorous inclusion criteria. For each study, NOS was applied to evaluate the quality of method, and we found all identified studies were deemed to be of high quality, except for one study . To ensuT. gondii can cause excessive expression of cytokines and chemokines as well as activation of astrocytes; these reactions might facilitate proinflammatory responses . ProinflSeveral limits should be considered in the interpretation of this review. First, our conclusions might be influenced since subjects in different regions usually had different genetic factors, environmental exposures, and lifestyles. Second, all studies in this review were based on case-controlled design; thus only association between exposure to T. gondii and PD risk could be investigated, not causal relationship. Third, potential language bias might exist in our review, as eligible studies were restricted to literature in English language. Fourth, subgroup analysis was not conducted due to limited information.This review does not suggest any association between T. gondii infection and development of PD. The pathogenic mechanisms of T. gondii in PD still remain incompletely clear; further researches are required to figure out the underlying mechanisms. Moreover, well-conducted and large cohort studies are warranted to further investigate the causal relationship between exposure to T. gondii and the development of PD. Whether treatment of T. gondii infection could effectively prevent or delay the progress of PD is also suggested to be tested in long-term and high-quality intervention studies.


CD14 polymorphisms are associated with an increased risk of cardiovascular events. So far, many studies have been conducted, whereas the results were not always consistent.Twenty-six articles involving thirty-seven datasets were recruited to evaluate the association between rs2569190 (9413 patients and 7337 controls), C-159T (4813 patients and 2852 controls) polymorphisms and cardiovascular diseases in a meta-analysis. The random or fixed effect models were used to evaluate the pooled odds ratios (ORs) and their corresponding 95% confidence intervals.p = 2.44 × 10− 5). Analysis after stratification by ethnicity indicated that rs2569190 was related to CVD in East Asian population  and a potential relationship in European . In the stratification of endpoints, the associations were found in CHD subgroup  and in AMI subgroup . However, we did not find any association between C-159T polymorphism with cardiovascular disease under any model.The strongest association was observed between rs2569190 and CVD in overall population  contains supplementary material, which is available to authorized users. Cardiovascular disease (CVD) is a major public health problem owing to associate increased risk of human mortality . By far The cluster of differentiation antigen 14 (CD14) is a lipopolysaccharide (LPS) receptor located on the surface of monocytes and macrophages and it is a multiple function inflammation cytokine which is mainly produced by mature mononuclear macrophage . CD14 isCD14 gene single nucleotide polymorphism (SNP) rs2569190, related to the translation start site at upstream of promoter region, was apparently relevant to atherosclerosis in the Czech population, the allele T of rs2569190 was also reported being a risk factoring for MI in European population. Then, a bunch of studies were carried out to verify the causal relationship between rs2569190 and CAD in diverse ethnics. Differently, the negative results that the polymorphism was not associated with CHD and MI were observed in European participants [It is generally accepted that genetic predisposition is a major risk factoring for atherosclerosis leading to CAD. In 1999, since Hubacek et al.  first reicipants . In addiicipants , 18. Meaicipants , 21. Besicipants , then thicipants . On the CD14 gene polymorphisms and CVD, the results were inconsistent. Population stratification might lead to inconsistent results, especially when allele frequency and incidence rate of the disease vary across ethnic groups. Meanwhile, inclusion of data that didn’t satisfy the requirement of meta-analysis would produce a spurious association. Therefore, in order to reduce the limitations of single study and to overcome the possible random errors, a large-scale meta-analysis involving multifarious ethnics was performed by us.Although many studies have been conducted so far to investigate the relationship between CD14 gene  and CVD, all published literature before December 2017 that researched the relationship between these polymorphisms and CVD risk were concluded. The electronic databases were used including PubMed databases , CNKI , and Web of Science were retrieved by using the keywords “CD14”, “C-260T”, “C-159T”, “rs2569190”, “polymorphism” connect to “CVD”, “coronary artery disease”, “coronary heart disease”, “myocardial infarction”, or “atherosclerosis” without language restrictions. Finally, we extracted data from the published articles, not included meetings or any conference abstracts. All of the included studies used either case-control or nested case-control design. Appropriate diagnosis criteria  and proper genotyping methods were used in most of the studies.To analyze the association between P value, (4) Relevant cardiovascular outcomes were angiographically confirmed  and coronary stenosis was diagnosed as at least one coronary artery stenosis (no less than 50% by a coronary angiography [Such major criteria must be followed for included studies (1) Original papers containing complete data, (2) Case–control or cohort studies that assessed the association of CD14 gene -159C/T and rs2569190 polymorphisms with CVD, and (3) Sufficient data to calculate the odds ratio (OR) or iography , 25), (5The primary reasons for excluded studies: (1) Case-only studies, review or meta-analysis articles, (2) Deviate from the major selection criteria, (3) Overlapping or that supplied inadequate data, (4) Repeated publications or the same authors employed similar data in different papers, the data was only used once.The study data was extracted based on standard protocol. For studies in which data could not be separated according to type of CVD from published data, cases were classified in the more inclusive category of coronary stenosis for the purposes of subsidiary analyses. Disagreement was settled by a consensus between all authors. Where essential information was not presented in articles, every effort was made to contact the authors. All procedures conformed to the guidelines for meta-analysis of observational studies in epidemiology. The following information were extracted independently by individual in our study: first author, year of publication, ethnicity, study design, types of CVD endpoints, HWE status among control, sample size of case and control, number of genotype and allele frequency.P value in heterogeneity test was higher than 0.1, the fixed effect model was used. Moreover, the random effect model was used [I2 = 100 %  × (Q − dƒ )/Q. The proportion of between-study variability attributable to heterogeneity was indicated by I2 value, and I2 values of 25, 50 and 75% were considered to be of low, moderate and high heterogeneity, respectively. If study groups revealed no heterogeneity, the similar results were produced in fixed and random effects models and, otherwise the random effects model usually produced wider CIs than the fixed effects model. In this meta-analysis, P value of less than 0.05 was considered a statistically significant.We calculated the allele frequency for each study in allele counting method, the Hardy–Weinberg equilibrium (HWE) was tested by using the Chi square test . We emplwas used –31. We uP value < 0.05 the statistical publication bias was considered [In order to get exacting search results, we evaluated possible publication bias by Egger’s linear regression text . If P vansidered . Moreovensidered . For sennsidered . StatistIn this meta-analysis, totally 218 relevant articles (145 for rs2569190 and 73 for C-159T) were searched. After reading titles and abstracts, we excluded irrelevant studies, 89 articles (54 for rs2569190 and 35 for C-159T) for further reading. Then, we excluded 57 articles (33 for rs2569190 and 24 for C-159T), because of no data, insufficient data, repeated date, family-based studies and not referring to cardiovascular disease (CVD). Thus, 32 articles (23 for rs2569190 and 9 for C-159T) met the study inclusion criteria. Lastly, after excluding three groups of data in which the control populations deviated from HWE , 35, 36 p = 0.001, I2 = 52.5%), but after stratified by ethnicity, the heterogeneity was resolved in any subgroup . An outstanding association was found in the overall population, under the random effect model , suggesting that the fixed effect model could be used.The association between C-159T and CVD was conducted in 13 independent studies with 7665 participants in this meta-analysis and all of datasets followed the inclusion criteria. The test of heterogeneity in the overall population was not significant  and C-159T (p = 0.077) under allele genetic model . It was verified that the T allele of rs2569190 apparently increased the risk of CVD, particularly in East Asian population. In stratified analysis by endpoints, rs2569190 was significantly associated with CHD, whereas, no any allele-specific or genotype-specific relation of the C-159T with cardiovascular disease was observed in diverse ethnicities.CD14 gene was considered modulating the capacity to stimulate inflammation through the regulation of CD14 gene expression and plasma soluble CD14 (sCD14) levels [The SNP rs2569190, locating in the promoter region of the ) levels , 58. In ) levels  and ecze) levels ; afterwa) levels . In 1999P = 6.15e-08, Blacks P = 0.04), meanwhile the same findings were also obtained by other investigators in another European population [P = 2.86 × 10− 8) as well as a certain relevance in the European, but no association in America population. Similarly, in the previous meta-analysis with 28 case–control articles testified a significant association between rs2569190 with CHD in East Asian, but they acquired a negative result in European populations under any genetic models [Recently, in the genome-wide association studies (GWAS) of Alex P et al. , a robuspulation . To checc models , which wc models , 21, 54,c models , 35, 36.P = 0.015), which was negative result in another meta-analysis study [In our meta-analysis with rs2569190, 16 studies for the European were included. After analyzed, the marginal significance was discovered in European subgroup , thrombus formation and myocardial infarction (MI) . Previoup = 0.05), recessive (p = 0.01) in the whole population with marginal significance, no significant association was found between them under dominant (p = 0.10). A significant association was detected in Chinese population (p < 0.05), while there was no significant association in the European subgroup (p > 0.05). To precisely explore the association between C-159T and CAD, we analyzed the data that were consistent with HWE. Ultimately, we carried out a comprehensive meta-analysis with 7665 participants. In contrast, we obtained the ineffective relationship between C-159T with any phenotypes of CVD in all subgroups from our study. Then, we compared to previous meta-analysis, and we found that they only included 5 articles about Chinese and 2 studies about European, such deficient data sets may lead to imprecise results. Simultaneously, we realized that the marginal significance under the allelic model and recessive model will be leaded by limited data sizes for C-159T, whereas our study included 13 researches to perform a more comprehensive study which will be more persuasive. In addition, the significant heterogeneity was easily found under all models in their study, of which the heterogeneity reached threshold (I2 = 83%) in the most relative recessive model. So many limitations were discovered, which did not emerge in our study, will produce contradictory conclusion.In 2015, Li et al.  performeAlthough no significant association between C-159T and CAD was found in our study, the positive results were discovered by previous studies . We founCD14 gene, located in 5q23–31, spans 3.9 kb, which encodes the glycoprotein with 375 amino acids [CD14 promoter at base pair − 260 from the major transcription start site and C-159T, in the 5′ flanking region of the CD14 gene at position − 159 [CD14 gene polymorphisms rs2569190 and C-159T has been reported being associated with a higher risk of CAD [CD14 gene promoter affect CD14 gene expression and lead to atherosclerosis, increase the risk of CAD, was not to be researched.In addition, no acids . The twoon − 159 –69. Manyon − 159 , 50 and k of CAD . Thus, tAlthough we revealed some new discoveries in this study, there were still several limitations should be taken into consideration. In our study, the overall sample size is large, but the size of each study is relatively small, the smallest sample is 54 cases and 70 controls , and we We conducted a meta-analysis to evaluate the effects of CD14 polymorphisms (rs2569190 and C-159T) on the risk of cardiovascular disease. This meta-analysis indicated that SNP rs2569190 significantly contribute to susceptibility and development of CVD, particularly in the East Asian population and in the subtype CHD group, in addithon, a potential association was observed in the AMI group, T allele acts as a risk factor for cardiovascular disease. However, we failed to acquire the positive association between rs2569190 and other subtypes of CVD. Meanwhile, the associations between C-159T polymorphism with CVD were not observed under any model. Further efforts should be put on investigating the association between the functional mutations within CD14 gene and CVD, and the interactions in potential gene-gene and gene-environment should be comprehensively analyzed.Additional file 1:Table S1. The information of publication bias in all Polymorphisms. Figure S1. Forest plot for the meta-analysis of the association between CD14 gene polymorphisms and CVD under the Dominant and Recessive models. Figure S2. Begg’s funnel plot of publication bias in the meta-analysis of the association of CD14 polymorphisms with CVD risk. Figure S3. Sensitivity analysis to assess the stability of the meta-analysis. (PDF 576 kb)


VEGFR2) polymorphisms and atherosclerotic cardiovascular diseases (ASCVD), with conflicting findings.Some previous studies explored associations between vascular endothelial growth factor receptor 2  and 95% confidence interval (CI) to estimate whether there are genetic associations between VEGFR2 rs1870377 (dominant comparison: 0.81 (0.73‐0.89), I2 = 56%; recessive comparison: 1.40 (1.25‐1.57), I2 = 34%; allele comparison: 0.81 (0.76‐0.87), I2 = 0%), rs2071559 (dominant comparison: 0.83 (0.76‐0.91), I2 = 0%; recessive comparison: 1.22 (1.07‐1.38), I2 = 0%; allele comparison: 0.86 (0.81‐0.92), I2 = 0%) and rs2305948 (dominant comparison: 0.79 (0.72‐0.87), I2 = 25%; recessive comparison: 1.44 (1.08‐1.92), I2 = 60%; allele comparison: 0.79 (0.68‐0.92), I2 = 73%) polymorphisms were all found to be significantly associated with susceptibility to ASCVD in general population. Subgroup analyses by type of disease revealed similar significant findings for rs1870377, rs2071559, and rs2305948 polymorphisms in coronary artery disease (CAD) subgroup. Besides, positive results were also found for rs1870377 polymorphism in ischemic stroke (IS) subgroup.Ten studies were included for this meta‐analysis (5474 cases and 8584 controls). VEGFR2 polymorphisms could be used to identify individual with elevated susceptibility to ASCVD.In summary, this meta‐analysis proved that these  Nevertheless, the findings of these studies were not always consistent and the sample size of each study was also statistically insufficient. In this meta‐analysis, we aimed to better analyze the relationship between VEGFR2 polymorphisms and ASCVD in a larger combined population.In recent years, some investigations already studied potential associations between 2This meta‐analysis was written in accordance with Preferred Reporting Items for Systematic Reviews and Meta‐analyses (PRISMA) checklist.2.1Eligible articles published before May 2019 were retrieved from PubMed, Web of Science, Embase, and CNKI by using the following key words:  AND (“polymorphism” OR “variant” OR “variation” OR “SNP” OR “mutation” OR “genome‐wide association study” OR “genetic association study”) AND . Additionally, we also checked the reference lists of all retrieved articles.VEGFR2 polymorphisms and ASCVD in human beings; (b) providing distributions of genotypes in cases and controls; (c) available full text in English or native language of the authors (Chinese). We excluded studies when more than one of the following conditions was met: (a) studies that were not about VEGFR2 polymorphisms and ASCVD; (b) reviews or comments; (c) case reports or case series. If we found repeated publications by the same authors, only the most comprehensive study was included for this meta‐analysis.Inclusion criteria for this meta‐analysis were as follows: (a) genetic association study about 2.2VEGFR2 polymorphisms in cases and controls. We also calculated the probability value  of Hardy–Weinberg equilibrium (HWE).Following information was extracted by two authors: the last name of the first author and publication year, country of the principal investigator and ethnicity of study participants, type of disease, total sample size of each study and the distribution of Newcastle–Ottawa scale (NOS) was used to evaluate the methodology quality of eligible studies.Data extraction and quality assessment were conducted by two authors independently. We wrote to the corresponding authors for extra information when we thought that important information was missed.2.3Z test to assess whether VEGFR2 polymorphisms were significantly associated with ASCVD, with the statistical significance P level set at .05. I2 statistics were used to evaluate between‐study heterogeneities. Random‐effect models (DerSimonian‐Laird method) were used if I2 exceeded 50%. Otherwise, meta‐analyses were conducted with fixed‐effect models (Mantel–Haenszel method). We also conducted subgroup analyses by type of disease. We tested the robustness of synthetic results in sensitivity analyses. We evaluated publication biases by funnel plots.Review Manager Version 5.3.3 was used in this meta‐analysis for statistical analyses. We used the 33.1Seventy‐two articles were identified by our comprehensive literature searching. Nineteen articles were retrieved for eligibility assessment after exclusion of irrelevant and duplicate articles. Another three reviews and five case series were subsequently excluded, and one study was excluded due to lack of essential data. Totally 10 eligible studies were ultimately included for this meta‐analysis , I2 = 56%; recessive comparison: 1.40 (1.25‐1.57), I2 = 34%; allele comparison: 0.81 (0.76‐0.87), I2 = 0%), rs2071559 (dominant comparison: 0.83 (0.76‐0.91), I2 = 0%; recessive comparison: 1.22 (1.07‐1.38), I2 = 0%; allele comparison: 0.86 (0.81‐0.92), I2 = 0%) and rs2305948 (dominant comparison: 0.79 (0.72‐0.87), I2 = 25%; recessive comparison: 1.44 (1.08‐1.92), I2 = 60%; allele comparison: 0.79 (0.68‐0.92), I2 = 73%) polymorphisms were all found to be significantly associated with susceptibility to ASCVD in general population. Subgroup analyses by type of disease revealed similar significant findings for rs1870377, rs2071559, and rs2305948 polymorphisms in CAD subgroup. Moreover, positive results were also found for rs1870377 polymorphism in IS subgroup  and rs2305948 (+1192G > A) polymorphisms were associated with reduced binding affinity of VEGFR2.VEGFR2 and VEGF, and ultimately influence individual susceptibility to ASCVD. Second, in stratified analyses, we noticed that the positive results were mainly driven by the CAD subgroup, which suggested that the magnitude of effects of VEGFR2 polymorphisms on individual susceptibility to CAD and IS might be somewhat different. However, given that the trends of associations in CAD and IS were similar, and the sample sizes of combined analyses with regard to IS were still relatively small. Maybe our meta‐analysis was still not statistically sufficient to detect the actual relationship between VEGFR2 polymorphisms and IS. So we call on further genetic association studies to confirm our findings, especially for VEGFR2 polymorphisms and IS. Third, the etiology of ASCVD is very complicated, consequently, we strongly recommend future studies to conduct haplotype analyses and investigate potential gene‐gene interactions to more comprehensively explore the effects of genetics on disease susceptibility.There are few points that should be considered when interpreting our meta‐analysis results. First, pre‐clinical studies proved that the minor allele of rs2071559 polymorphism (−604 T > C) could lead to decreased transcription activity of 1870377 +19A > T aVEGFR2 polymorphisms and ASCVD. Regrettably, most of included studies only focus on genetic associations, so we could not conduct analyses regarding genetic‐environmental interactions.VEGFR2 polymorphisms. Since no any other VEGFR2 polymorphisms were studied by at least two eligible studies with regard to their associations with ASCVD, this meta‐analysis only focus on relationship between three common VEGFR2  polymorphisms and ASCVD.This meta‐analysis has some limitations. First, our meta‐analysis results were derived from unadjusted combined analyses, and failure to adjust for some crucial variables may impact the precision of our findings.5VEGFR2 rs1870377, rs2071559, and rs2305948 polymorphisms could be used to identify individual with elevated susceptibility to CAD. Moreover, rs1870377 polymorphism could be used to identify individual with elevated susceptibility to IS. However, further studies with larger sample sizes still need to verify our findings, especially for IS.In summary, this meta‐analysis proved that The authors declare no potential conflict of interests.Li Wang and Bing Wang conceived and designed this study. Li Wang and Hui Ge conducted the systematic literature review. Longyun Peng performed data analyses. Li Wang and Bing Wang drafted the manuscript. All authors gave final approval and agree to be accountable for all aspects of work ensuring integrity and accuracy.This article does not contain any studies with human participants or animals performed by any of the authors, thus no ethical approval is required.FIGURE S1 Funnel plotsClick here for additional data file.


Previous studies have potentially linked PGC-1α rs8192678 polymorphism to type 2 diabetes mellitus (T2DM) risk, but the results remain inconsistent in different populations and are not conclusive. We developed a new meta-analysis approach to systematically identify the association between PGC-1α rs8192678 polymorphism and T2DM, and we have comprehensively assessed different ethnic groups to validate our findings.  Genome-wide association study (GWAS) provides an unprecedented opportunity to reveal substantial genetic contribution to type 2 diabetes mellitus (T2DM) and glycemic identification of allelic heterogeneity and population-specific genetic variants, yet it also faces difficulty due to the vast amount of potential confounding factors and limited availability of clinical data. To identify responsible susceptibility loci and genomic polymorphism for T2DM and glycemic traits, we have systematically investigated a genome-wide association study related to T2DM. Although GWAS has captured many common genetic variations, which are related to T2DM, each risk allele (RA) of single-nucleotide polymorphisms (SNPs) at these loci is not conclusive. Therefore, it is common to present a combination of several SNPs to infer T2DM risk, yet it is still insufficient to be deterministic. To streamline the identification of a deterministic genetic variation in T2DM, we developed this meta-analysis as a showcase to comprehensively identify the association between cumulative RAs and T2DM risk by combining different studies in reported literature and databases. After all, we identified that PGC-1 We performed comprehensive information retrieval and knowledge discovery meta-analysis by searching extensively published literature and different electronic databases to acquire eligible studies for the above association study. We developed a method to use pooled odds ratios (ORs) and their corresponding 95% confidence intervals (CIs) in five genetic models  to identify the relationship among ethnicity subgroup analyses comprehensively. α rs8192678 polymorphisms of all subjects showed a significant association with T2DM susceptibility under all genetic models: allelic , dominant , recessive , homozygous , and heterozygous . In the subgroup analysis, we identified a significant association between PGC-1α rs8192678 polymorphism and T2DM in the Caucasian and Indian populations under all genetic models we investigated. This is the most comprehensive study of the subject to date.  We identified 20 eligible studies consisting of 16,182 subjects  in our meta-analysis. PGC-1 Our development of meta-analysis revealed that the minor allele (A) carriers, especially AA genotype carriers, can lead to risk of T2DM in the Caucasian and Indian populations. This is the first report that such risk has been confirmed. Our finding shed new light into the genetic alteration in T2DM. T2DM is a significant global health issue with potential life-threatening complications if not controlled well. Over 451 million people worldwide were estimated to live with diabetes according to the International Diabetes Federation . A largeγ coactivator-1α (PGC-1α) is crucially important to T2DM. PGC-1α is a multifunctional regulatory factor originally identified as a coactivator of PPARγ in 1998 [α, located on chromosome 4p15.1, is highly expressed in the liver and skeletal muscle and is involved in maintaining glucose, lipid, and energy homeostasis. PGC-1α, as a crucial gene regulatory element in various metabolic processes, has been shown to play pivotal roles in the development of obesity, insulin resistance, and T2DM [α, such as rs8192678G>A, rs2970847C>T, rs3736265G>A, and rs3755863 C>T, with risk of T2DM were studied [α gene has been speculated in T2DM risk in some studies [After comprehensive investigations on T2DM, we found that peroxisome proliferator-activated receptor  in 1998 . PGC-1α,and T2DM . In the  studied , while a studies –14; suchα rs8192678 (G>A) polymorphism and T2DM was extensively investigated more comprehensively by combining relevant studies. In addition, we combined ethnicity in all eligible studies to elucidate the explicit association of PGC-1α rs8192678 (G>A) polymorphism and T2DM.In this study, the association of PGC-1The electronic databases, both in English and Chinese, including PubMed, Springer, Chinese National Knowledge Infrastructure, SinoMed, and Wanfang were searched for collecting relevant literatures published from January 2001 to May 2018. For all databases, we expanded broad search terms such as “PPARGC1A,” “PGC-1alpha”, “T2DM,” “diabetic,” “insulin,” “hyperglycemia,” “polymorphism,” and “mutation” for more comprehensive information harvest. We do not set up any other filter to ensure all relevant information will be included in our investigations.Studies published in peer-reviewed journalsα polymorphism and T2DM riskCase-control studies referring to the association between PGC-1Cases and controls with allele and genotype frequenciesP > 0.05).Genetic variants of controls that met the Hardy-Weinberg equilibrium  independently reviewed full texts to carefully select eligible papers and extracted all information, including the first author, year of publication, ethnicity of subjects, age, gender, BMI, diagnostic criteria, sample size, molecular methods for genotype testing, and allele and genotype distribution in cases and controls. The third researcher (Wen-jia Peng) reviewed, cross-verified, and then finalized the carefully selected literature. Any disagreements were resolved by all 3 researchers through careful reexaminations altogether. If similar data were reported more than two times in different papers, only one was adopted in order to eliminate any bias. The genotype frequencies of controls in all included literatures were verified by HWE.The Newcastle-Ottawa Scale (NOS)  tool wasα gene. The pooled odds ratio (OR) with a 95% confidence interval (95% CI) was calculated in different genetic models, allelic (A vs. G), dominant (AG + AA vs. GG), recessive (AA vs. AG + GG), homozygous (AA vs. GG), and heterozygous (AG vs. GG) genetic models. Subgroup analysis by ethnicity was also conducted. Heterogeneity across studies was assessed via a chi-square test and I2. The fixed effects model was used for no heterogeneity , whereas the random effects model was selected. The HWE was examined via the chi-square test. A funnel plot was used to detect potential publication bias, and the funnel plot symmetry was evaluated using Egger's linear regression testing on the OR. If publication bias was indicated, we further evaluated the number of missing studies by the Trim and Fill method and recalculated the pooled risk estimation with the addition of those missing studies [All statistical analyses were performed using STATA version 12.0 and RevMan v5.3 (the Cochrane Collaboration) software. The G allele is wild, and the A allele is mutated for the rs8192678 G>A polymorphism in the PGC-1 studies .After duplicates were removed, 68 papers of potentially relevant studies were included initially. Then, 32 papers were excluded by reading abstracts carefully. Among those remaining relevant studies, 16 papers  were also excluded after reading the full texts . Thus, 2α rs8192678 polymorphisms have shown a significant association with T2DM risk under allelic , dominant , recessive , homozygous , and heterozygous  genetic models , recessive , and homozygous  genetic models , dominant , recessive , homozygous , and heterozygous  genetic models .This significant association was also found in the Indian population under allelic , dominant , recessive , homozygous , and heterozygous  genetic models .No AA genotype carrier was found in the African population, nor was there any significant association detected in this study.T = 3.22, P = 0.004) under the recessive genetic model. The Trim and Fill method was used to recalculate the pooled risk estimate. There was some indication of asymmetry (seven studies trimmed), but the results were still stable before and after this analysis  based GWAS approaches. However, up to today, the quantitative association of cumulative risk alleles (RAs) of such SNPs with T2DM risk has not been deterministic, limiting the potential applications of this type of research. Hence, the aim of this study was to identify deterministic genetic alteration associated with T2DM. The comprehensive meta-analysis of cross-ethnicity studies revealed T2DM risk in relation to PGC-1α gene rs8192678 polymorphism and T2DM susceptibility was detected under different genetic models. Individuals with A allele carriers can have an increased risk of T2DM, especially among the Caucasian and Indian populations. After subgroup analysis, a significant association was detected in the Caucasian population under all genetic models, and also found in the Indian population. Furthermore, people with AA genotype carriers can have a higher risk of T2DM compared to people with other genotype carriers. The association was detected in the East Asian population under recessive and homozygous genetic models. No significant association was found among the Africans because of the only one study with small sample size and without AA genotype carriers in control. Heterogeneity showed reduction after ethnicity subgroup analysis. This infers that ethnicity can be a main reason for heterogeneity. The possible publication bias, as detected by Funnel plot and Egger's linear regression test, suggests that further studies are still needed to validate the conclusions, while the sensitivity analysis supports that our results are stable, robust, and reliable.In this meta-analysis, the association between PGC-1α rs8192678 polymorphism and T2DM. The association between PGC-1α polymorphism and T2DM, insulin secretion, and other related indicators was also explored in different countries, but with inconsistent and contradictory conclusions [α rs8192678 polymorphism and T2DM in the Caucasian population. Eight Caucasian studies from 2001 to 2005 were included in their meta-analysis, and only 2 genetic models  were adopted. Because both unrelated and genealogical samples were included in their analysis, the comparability of studies might be decreased and their conclusion may be more limited, compared to the results from our study. In 2011, Yang et al. [α rs8192678 polymorphism and T2DM but only included 8 studies. However, a high level of heterogeneity among these studies was identified with an I2 value of 81.58%. Hence, the reliability of the previous results has been questionable. We performed a much more comprehensive investigation, and in our meta-analysis, all included studies were detected by HWE. The selection of literature by inclusion and exclusion criteria was strictly followed to reduce the potential selection bias. We performed more comprehensively five genetic models with subgroup analysis stratified by ethnicity and literatures (after 2010). Therefore, our much more comprehensive approach offers more reliable results with new updates.Initially, Ek et al.  performeclusions –23. The clusions  found a g et al.  also perg et al.  investigDespite our promising results, several potential limitations should also be addressed. Firstly, T2DM is a complex multifactorial disease, and we only considered the individual polymorphism without taking into account the interaction with other polymorphisms or environmental factors . Secondly, the study did not consider other relevant factors , due to the limitation of availability of original research data. Thirdly, obesity is an important intermediate factor in the development of T2DM, but the definitions of obesity were different or not available in our included studies. It would be valuable and interesting to include obesity in the association with subgroup analysis. Finally, sample sizes were insufficient in the Indian and African populations. However, even though those factors are obviously related, none of them is indispensible in T2DM development, since T2DM can occur without any of the above factors. Therefore, despite of those limitations, a reliable conclusion can be still made from our comprehensive study.α rs8192678 polymorphism and T2DM susceptibility has been detected in this meta-analysis. Individuals with A allele, especially AA genotype carriers, can be more susceptible to T2DM in the Caucasian and Indian populations. Such association in East Asian and African populations needs to be further explored in larger sample studies. Further research about the interaction effect on PGC-1α rs8192678 polymorphism with other SNPs or environmental factors will be followed. Our finding revealed an explicit genetic variation in T2DM development. Our study provides further insight into the mechanisms by which genetic variation influences type 2 diabetes risk and glycemic traits, and it further supports that the genetic variant for type 2 diabetes risk can confer the risk diversity among different ethnicities.A significant association between PGC-1