Cleft lip and palate transmembrane 1-like (CLPTM1L) gene rs402710 (C > T) and rs401681 (C > T) polymorphisms have been widely studied for their potential relation to cancer risk, but studies have produced conflicting results. To systematically evaluate the association between these two polymorphisms and overall cancer risk, we conducted a comprehensive meta-analysis on all relevant articles found in the PubMed and EMBASE databases published prior to May 1, 2017. There were 26 articles with 28 studies, including 30,770 cases and 34,089 controls, for the rs402710 polymorphism and 38 articles with 48 studies, including 67,849 cases and 328,226 controls, for the rs401681 polymorphism. The pooled results indicated that both rs402710 and rs401681 polymorphisms are significantly associated with decreased overall cancer risk. In our stratification analysis, a significant association of the rs402710 polymorphism with lung and bladder cancers was identified among Asian and Caucasian populations in both hospital-based and population-based studies. The rs401681 polymorphism was significantly associated with a decreased risk of lung cancer, bladder cancer, and basal cell carcinoma in Asians and in hospital-based studies. CLPTM1L gene rs402710 and rs401681 polymorphisms thus have a protective association with various types of cancer, especially lung cancer among Asians. The number of people who will develop and die from cancer is expected to climb rapidly around the world . In 2016Telomerase reverse transcriptase (TERT) and cleft lip and palate transmembrane 1-like (CLPTM1L) gene variation on cancer susceptibility. The TERT gene encodes the rate-limiting catalytic subunit of the telomerase enzyme that is vital to the maintenance of telomere DNA length, chromosomal stability, and cellular immortality [CLPTM1L, encodes a protein linked to cisplatin resistance and associated with the susceptibility to cleft lip palate. CLPTM1L was found to be overexpressed in cisplatin-resistant ovarian cancer cells and to promote apoptosis in cisplatin-sensitive cells [TERT-CLPTM1L genetic variants, including single nucleotide polymorphisms, have been reported by several genome-wide association studies (GWAS) to associate with the risk of multiple cancer types [CLPTM1L rs402710 and rs401681 polymorphisms and overall cancer risk.Recently, a number of investigations have been conducted to study the potential influence of ortality . CLPTM1Lve cells . TERT-CLer types –11. Amoner types , but other types , 14. Forer types , but theer types , 17. We As shown in Figure The genotypic distribution of rs402710 and rs401681 polymorphisms in the controls in almost all of the studies followed Hardy-Weinberg equilibrium (HWE); four did not , except for the heterozygous model (p = 0.121). As to the rs401681 polymorphism, we also observed considerable heterogeneities . So, the random-effects model was applied to generate crude ORs and 95% CIs. In addition, the leave-one-out sensitivity analysis indicated that no single study altered the corresponding pooled ORs and 95% CIs.Substantial heterogeneities were observed among all studies of the rs402710 polymorphism and cancer risk  and rs401681 polymorphism .The shape of our funnel plot did not reveal any evidence of obvious asymmetry for the rs402710 polymorphism  and rs401681 (C>T) polymorphisms and overall cancer risk. In this meta-analysis, based on the database of publications in respect of the rs402710 polymorphism and the rs401681 polymorphism, among which there are 26 articles with 28 studies including 30,770 cases and 34,089 controls for rs402710 polymorphism and 38 articles with 48 studies including 67,849 cases and 328,226 controls for rs401681 polymorphism respectively, we observed both rs402710 and rs401681 polymorphisms were significantly associated with a decreased overall cancer risk. In the stratification analysis, as to the rs402710 polymorphism, a statistically significant association was also identified for lung and bladder cancer. Likewise, the rs401681 polymorphism was significantly associated with a decreased risk for lung cancer, bladder cancer and basal cell carcinoma. Interestingly, an increased risk of melanoma and pancreatic cancer was found for the rs401681 polymorphism.To the best of our knowledge, the meta-analysis in this study is the largest to study the associations between Up until now, there had been only one meta-analysis (in 2013) focusing on rs402710 and rs401681 polymorphisms and overall cancer risk ; which oSeveral limitations in our meta-analysis should be acknowledged. First, we only included the studies published in English, which may have missed publications in other languages. Second, in the stratification analysis by cancer type and ethnicity, the sample sizes are relatively small and may be insufficient, which may have diminished the statistical power. Third, due to the lack of raw information such as age, sex, body mass index, smoking habits, and alcohol consumption, this meta-analysis was conducted on the basis of unadjusted estimates. Therefore, larger and well-designed studies with different types of cancer and multiple ethnic with sufficient raw information are warranted to validate the findings of this study.CLPTM1L gene rs402710 and rs401681 polymorphisms were associated with a decreased risk of various types of cancer, especially for lung cancer among Asians.In conclusion, the meta-analysis in this study demonstrated that CLPTM1L polymorphisms and overall cancer risk published prior to May 1, 2017. The following search terms were used: “CLPTM1L”, “5p15”, “polymorphism”, “variant”, “cancer”, “tumor” and “carcinoma”. Additional relevant studies were also retrieved manually from all the relevant publications, including the eligible original publications and reviews. Only the latest or the largest study would be included in the final meta-analysis.PubMed and EMBASE electronic databases were screened for publications investigating the association between CLPTM1L rs402710 and rs401681 polymorphisms and the risk of all types of cancer; (2) case-control or cohort studies; (3) published in English; (4) including sufficient data for genotype frequencies; and (5) providing sufficient data to calculate ORs and 95% CIs. Exclusion criteria include: (1) insufficient raw data; (2) meta-analyses, reviews and comments; (3) duplicates of previous publications; and (4) case-only studies. Studies that were departure from HWE in controls were also excluded, unless further evidence indicated that another polymorphism was in agreement with HWE.All studies included in the current meta-analysis met the following criteria: (1) evaluating the association between Two investigators extracted the following information from all the eligible studies independently: surname of the first author, publication year, country of origin, ethnicity, cancer type, control source, genotyping method, number of cases and controls, and genotype frequencies of cases and controls. Different ethnicities were stratified into Asian, African, and Caucasian. The source of controls was categorized as hospital-based and population-based studies. Any disagreements were discussed and resolved until the consensus was reached.CLPTM1L rs402710 and rs401681 polymorphisms and the risk of all types of cancer were assessed by the calculation of crude ORs and 95% CIs. Stratification analyses were further performed according to ethnicity, cancer type and control source. The pooled ORs and 95% CIs were estimated for rs402710 polymorphism under the homozygous (TT vs. CC), heterozygous (CT vs. CC), recessive (TT vs. CT+CC), and dominant (CT+TT vs. CC), as well as allele comparison model (T vs. C) and so was rs401681 polymorphism under the homozygous (TT vs. CC), heterozygous (CT vs. CC), recessive (TT vs. CT+CC), and dominant (CT+TT vs. CC), as well as allele comparison model (T vs. C).The strength of association between the p > 0.10), the fixed-effects model (the Mantel-Haensze method) was used [2 test [The Chi square-based Q-test was applied to evaluate between-study heterogeneity. If the studies were found to have no heterogeneity  and were two-sided, with


The present meta-analysis was intended to explore the relationship between the X-ray repair cross complementing 1 (XRCC1) polymorphisms  and cervical cancer risk. Several electronic databases were searched systematically and bibliographies of relevant papers were identified carefully. Then, a meta-analysis was performed based on eligible studies in various genetic models. Pooled odds ratios (OR) with 95% confidence intervals (95% CI) were employed to evaluate the strength of associations between the XRCC1 polymorphisms and cervical cancer risk. Additionally, heterogeneity analysis and sensitivity analysis were done if necessary. Totally, 11 articles involving 2092 cases and 2803 controls were included. Taken together, there was no obvious association between the Arg194Trp or Arg280His polymorphism and cervical cancer risk. Considering the great heterogeneity, subgroup analysis was done, but the pooled result remained stable. Nevertheless, the association between the Arg399Gln polymorphism and cervical cancer risk showed distinct statistic significance in the allele model, dominant model, homozygous model and heterozygous model. In view of the exiting heterogeneity, we did subgroup analysis stratified by ethnicity, resulting in the fact that the Arg399Gln polymorphism was related to the decreased risk of cervical cancer. The Begg's test and Egger's test were used to find no publication bias. To conclude, the current meta-analysis indicated that the XRCC1 Arg399Gln polymorphism decreased the risk of cervical cancer, while the Arg194Trp and Arg280His polymorphisms were not associated with cervical caner risk. Certainly, a well-designed large-scale multicenter study is warranted to confirm the finding. Nowsdays, cervical cancer is one of the most common genital tract carcinomas and has become a challenging health issue confronted by women throughout the world. It seriously threatens women's quality of life arising from reproductive endocrine function's damage caused by this malignancy, and brings about great morbidity and economic burden. Infection with high-risk types of human papillomavirus (HPV) is the main causative factor for developing cervical intraepithelial neoplasia (CIN) which is a precursor lesion for cervical cancer. While, not all women who are infected with HPV will certainly progress into cervical cancer, suggesting that there are still other factors playing a role in the pathogenesis of cervical cancer. For example, ultraviolet, ionizing radiation and environmental chemical agents can lead to DNA damage, initiating certain human cancers –9.http://egp.gs.washington.edu). And eventually these variants alter XRCC1 function.In human body, DNA repair genes are considerable factors in the prevention of genomic injury and sequential carcinogenesis. So variants of DNA repair genes might be able to impair DNA repair ability and have been suggested to be associated with cancer risk. X-ray repair cross complementing group 1 ( XRCC1) gene is a typical DNA repair gene. It is located at chromosome 19q13.2-13.3 and encodes the scaffolding protein . The proA great many epidemiologic studies have been conducted to evaluate the role of the XRCC1 polymorphisms  on cervical cancer risk –24. But We initially retrieved 46 articles through various electronic databases. After removing reviews, meta-analysises, basic experimental studies, we got 32 articles needing screening the full-text. While 20 articles did not present available data and one was a duplicate study . Consequn = 133).  examination results of the included studies and the XRCC1 polymorphisms genotype distribution in cases and controls were displayed in Table P > 0.05). considering the moderate to great heterogeneity among studies, we performed subgroup analysis stratified by the degree of cervical lesion. However, the finding that the pooled OR still incorporated 1.0 showed that the Arg194Trp polymorphism had no association with the risk of cervical cancer. Then we excluded three studies which were not consistent with HWE [For XRCC1 Arg194Trp polymorphism, there were seven studies, involving 1315 cases and 1633 controls, evaluating the connection between it and cervical cancer susceptibility. All the studies were done among the Asian population apart from one study . Overallwith HWE , 22, 23 P > 0.05). However, the heterogeneity among studies droped to zero when excluding the study which didn't conform to HWE. Despite of this, the pooled results stayed stable when we eliminated the one [With regard to XRCC1 Arg280His polymorphism, four articles including 2015 objects (784 cases and 1231 controls) offered data about the association between it and cervical cancer risk. On the whole, the heterogeneity among studies were quite huge, the random model was employed to weigh the strength of the association. While the remarkable link between this genetic locus and cervical cancer wasn't witnessed in all models (P < 0.00001), dominant model , homozygous model , heterozygous model . . All the stuies were in accordance with HWE except one study [P = 0.00001; GlnGln + ArgGln vs. ArgArg: OR=0.06, 95% CI = 0.03–0.12, P < 0.00001; GlnGln vs. ArgGln + ArgArg: OR = 0.70, 95% CI = 0.61–0.81, P < 0.00001; GlnGln vs. ArgArg: OR = 0.44, 95% CI = 0.28–0.68, P = 0.002; ArgGln vs. ArgArg: OR = 0.54, 95% CI = 0.40- 0.72, P < 0.0001) and the Caucasian . , implying that the ethnicity exerted no influence on the association between the XRCC1 Arg399Gln polymorphism and the risk of cervical cancer.Considering the great heterogeneity among studies, the random-effect model was applied and subgroup analysis stratified by ethnicity Figure  and the P < 0.05), yet the final results were not substantially altered after excluding those. Simultaneously, the studies with quite large or small sample sizes were deleted one by one in order to test the stability of pooled results. Moreover, sequential deletion of each study was utilized to perform sensitivity analysis in all models. However, the pooled ORs did not show quantitative changes when excluding any study, suggesting that the results of this meta-analysis were stable and reliable. Sensitivity analysis of the association between the XRCC1 Arg399Gln polymorphism and the risk of cervical cancer in homozygous model was showed Figure Although some studies wasn't consistent with the balance of HWE in control groups (P > 0.05).The Begg's test and Egger's test were done in all models showing that there was no statistical evidence for publication bias. Publication bias of the XRCC1 Arg399Gln polymorphism in homozygous model was shown in Figure Cervical cancer is still the second most common malignant tumor among women and heavily threatens women's health in the world. To improve this embarrassing situation, risk factors concerning cervical cancer should be indentified timely and controlled effectively. There have exited several case-control studies focusing on the relationship between individual susceptibility or genetic variants and cervical cancer , 24, 26.Certainly, there have emerged several other meta-analysises concerning the link between XRCC1 polymorphisms and cervical cancer risk. A latest meta-analysis exploring the association between the Arg399Gln polymorphism and cervical cancer showed that the Arg399Gln polymorphism increased the risk of cervical cancer . The resLikewise, some limitations of this meta-analysis should be mentioned even though considerable effort and resources have been put into testing the possible association between the XRCC1 polymorphism and cervical cancer risk. On the one hand, we retrieved relevant articles only through electronic databases, leading to a potential bias caused by the lack of unpublished articles which would not be available in the electronic databases. On the other, although the great heterogeneity among studies had no effect on the pooled result, yet the heterogeneity could not be neglected completely.To conclude, the current meta-analysis indicated that the XRCC1 Arg399Gln polymorphism decreased the risk of cervical cancer, while the Arg194Trp and Arg280His polymorphisms were not associated with cervical caner risk. Certainly, to further evaluate the association between XRCC1 polymorphisms and cervical cancer susceptibility, a well-designed large-scale multicenter study is warranted to confirm the finding.A systematic literature search was done through PubMed, Web of Science, EMBASE and the Cochrane Library up to July 2016 in English. The search terms included “X-ray repair cross complementing protein 1”, “XRCC1”, “Arg194Trp”, “rs1799782”, “Arg280His”, “rs25489”, “Arg399Gln” or “rs25487”; “poly-morphism”, “variant”, “genotype”, “polymorphism” or “SNP”; “cervical” or “cervix”; “cancer”, “carcinoma”, “neoplasm”, “tumor” or “ malignancy “and the combinations. Besides, the relevant references of identified studies were screened carefully for potential articles.The included studies have to meet the following criteria: 1) investigating the association between XRCC1 polymorphisms  and risk of cervical cancer; 2) studies on human beings; 3) genotype frequencies were available both in case and control groups; 4) subjects in control groups should have no cancer history, previous radiotherapy and chemotherapy history and a family history of tumor; 5) the diagnosis of the cases was based on pathology. The study with the following criteria was excluded: 1) abstracts, case reports, letters, comments, editorials, reviews and mata-analysises; 2) studies lacking relevant data. What's more, the most recent study was included once the studies were duplicated. Any one study was screened by two authors independently and disagreements were resolved by discussing with a third author.Two investigators simutaneously extracted characteristics of the included studies according to the inclusion and exclusion criteria and the results were checked by a third reviewer. The data extracted from each study included first author, year of publication, country of origin, ethnicity, source of the control group, genotyping method and numbers of case and control subjects. Ethnicity was classified as ‘‘Caucasian’’, ‘‘Asian’’ and ‘‘Mixed’’.The methodological quality assessment was performed based on the modified scoring system used for studies in genetic epidemiological issues.  Points wP < 0.05 was considered statistically significant. To test the heterogeneity among studies, we assumed the I2 and Q statistic. We adopted random effect model if there was great heterogeneity (I2 greater than 50%). Otherwise, we adopted the fixed effect model. At the same time, we conducted a subgroup analysis according to ethnicities. To assess the stability of the finding, we performed sensitivity analysis. Each study involved in this meta-analysis was deleted respectively to reflect the influence of the individual data exerted on the pooled OR. HWE of the genotype frequencies in the control group of each study was assessed by χ2 test and P > 0.05 was considered to be consistent with HWE [Review Manage version 5.2.0  and STATA version 11.0 software  were applied to carry out statistical analysis. The association between XRCC1 polymorphisms and cervical cancer risk was estimated in the allele model, the dominant model, the recessive model, the homozygous genetic model and the heterozygous genetic model. To evaluate the strength of associations, the summary odds ratio (OR) and 95% confidence interval (CI) were calculated through fixed/random effects mode. with HWE . For thewith HWE , 34.


TGFB1 gene polymorphisms (SNP rs1800470 and rs1800471) may be associated with the risk of CAD. In this meta-analysis, the relationship between these two variations and the risk of CAD was explored.Epidemiological studies have investigated the role of transforming growth factor-β1 (TGF-β1) in chronic allograft dysfunction (CAD) following kidney transplantation. MEDLINE, EMBASE, the Cochrane Central Register of Controlled Trials (CENTRAL), Embase, the Chinese CNKI and WANFANG databases were searched. Data were extracted and pooled results were estimated from odds ratios (ORs) with 95% confidential intervals (95% CIs). Quality assessments were performed, and publication bias of all eligible studies examined.TGFB1 haplotypes between the CAD recipients and control subjects when “HIGH” with “INTERMEDIATE” and “LOW”  were compared. No significant association was found between the TGFB1 codon 10 or codon 25 and the CAD risk in all five genetic models.Eight studies with 1038 subjects were included in our analysis. According to the effects on TGF-β1 secretion, haplotypes were categorized as “HIGH”, “INTERMEDIATE” and “LOW”. The combined results showed a statistically significant difference of TGFB1 codon 10/25 T/T G/G and T/C G/G genotypes, associated with increased production of TGF-β1, was linked with CAD risk following kidney transplantation. Moreover, no significant difference was found between TGFB1 codon 10 or codon 25 and the development of CAD.Our meta-analysis has found the haplotype of  Kidney transplantation is the optimal therapy for end-stage renal disease . Short-tCAD is mainly characterized as glomerular sclerosis, tubular atrophy and interstitial fibrosis . VariousTGFB1 is mapped on to chromosome 19q13.1–13.3 with seven exons and six introns, and the expression of TGFB1 and production of TGF-β1 is related to single nucleotide polymorphisms (SNPs). TGFB1 contains two SNPs, +869T/C at codon 10 (rs1800470) and +915G/C at codon 25 (rs1800471) contributing to variation in TGF-β1 production both in vitro and in serum [TGFB1 SNPs, and CAD has been explored in several studies [TGF-β1 is a multifunctional growth cytokine in humans and one of the significant mediators of wound healing and tissue regeneration . TGFB1 iin serum . Since f studies . HoweverBased on the crucial role of TGF-β1 in CAD pathogenesis, we performed meta-analysis to investigate the contributions of two TGFB1 SNPs to CAD risk.A comprehensive literature search was performed in PubMed, the Cochrane Central Register of Controlled Trials (CENTRAL), Embase, the Chinese CNKI and WANFANG databases (updated on December 20st 2016) by two independent authors (Kun L and Xuzhong L). The following keywords were used: (transforming growth factor-beta 1 OR TGFB1), AND (polymorphisms OR SNPs OR variants), AND , AND . The equivalent Chinese terms were used in the Chinese databases. Furthermore, the reference lists of all studies included in the meta-analysis were also reviewed for possible inclusion.TGFB1 SNPs and chronic allograft dysfunction after kidney transplantation; (2) available information on the frequencies of genotype or allele in case and control groups; (3) all subjects from three allelic groups were derived from a population within the same geographic area and ethnic background; (4) full-text article were published in English or Chinese. The exclusion criteria were: (1) Case reports; (2) reviews; (3) animal experiment, chemistry, or cell line studies; (4) studies in a language other than English or Chinese; (5). No eligible or insufficient data frequencies of genotypes for each polymorphism could be extracted for meta-analysis. Two authors (Kun L and Xuzhong L) independently assessed and selected trials for final analysis with discrepancies resolved by consensus.Studies were included if they met the following eligibility criteria: (1) case-control studies designed to investigate the relationship between TGFB1 SNPs were collected using a standardized data extraction form. Missing data was examined by contacting the first or corresponding author.Two investigators (Kun L and Xuzhong L) independently extracted relevant data from all selected studies and reached consensus for all items. Basic characteristics of patients were collected as following: first author’s name, year of publication, nation/race, number of subjects, the proportion of male subjects, mean age and method of genotype. In addition, the genotype and allele frequencies of The methodological quality of each included study was assessed by the Newcastle-Ottawa quality assessment scale (NOS)(18). Each study was evaluated on the standard criteria and categorized based on three factors, including selection, comparability and exposure. Scores ranged from 0 to 9; 9 points represented the highest quality and lowest risk of bias.TGFB1 polymorphisms and CAD using odds ratio (OR) with 95% confidence intervals (95% CIs) in a dominant model, a recessive model, a co-dominant model, a co-recessive model and an allele model. A p value less than 0.05 was considered statistically significant. Heterogeneity test among trials was determined by I2, and was defined as 100% *(Q-df )/Q, where Q is Cochran’s heterogeneity statistic and df is the degrees of freedom, with a fixed-effect model set at low statistical inconsistency (I2 < 25%). Otherwise, a random-effects model was selected, which is better adapted to clinical and statistical variations [The pooled data was used to assess the strength of the association between riations –19. To eA flow diagram of the screening process for included studies is shown in Figure TGFB1 codon 10 and codon 25. Four studies [Basic characteristics of eligible studies are shown in Table  studies , 20–22 i studies , 3 Europ studies , 17, 24 TGFB1 polymorphisms and CAD. For TGFB1 codon 10 T/C, no significant difference was found between CAD recipients and control subjects in all five models  found that secretion of the proline form of TBG-β1 was 2.8 times that of the leucine form of TGF-β1 at codon 10 in transfected Hela cells, therefore the proline mutation at codon 10 could increase the amount of TGF-β1 protein secretion. Similarly, Cambien et al. [TGFB1 codon 10 or codon 25 with the CAD risk was found in all five models, suggesting that the single mutation in codon 10 and codon 25 were not responsible for the pathogenesis of CAD after kidney transplantation.The n et al.  believedTGFB1 gene, and except for codon 10 and codon 25, rare studies have focused on other SNPs, which may play a role in the pathogenesis of CAD.Notably, these results should be interpreted with caution. The case number of recipients with CAD included in our analysis is relatively small, which could lead to relatively high heterogeneity. Due to the limited number of studies in our analysis, we categorized CAN, confirmed by allograft renal biopsy, into the diagnosis of CAD, instead of subgroup analysis. Finally, there were at least 10 SNPs reported in the TGFB1 gene codon 10/25 T/T G/G and T/C G/G genotypes, associated with increased production of TGF-β1, was linked with CAD risk following kidney transplantation. Moreover, no significant difference was found between TGFB1 codon 10 or codon 25 and the development of CAD. Further studies incorporating subjects with difference ethnic backgrounds combined with re-sequencing of the marked region and functional evaluations are warranted.In conclusion, our meta-analysis found the haplotype of


ADPN I148M polymorphism has been consistently reported to play a role in liver-associated diseases, such as alcoholic liver disease, chronic hepatitis C, and liver fat and fibrosis in nonalcoholic fatty liver disease. This significant association was also indicated in a series of hepatocellular carcinoma (HCC) studies, where the significance may be affected due to the small sample sizes. The aim of this study was to reexamine the ADPN-HCC association by use of meta-analysis. Biweekly computer-based literature searches plus manual screening were undertaken in an effort to identify all studies that met the predefined inclusion criteria. The Mantel–Haenszel method was selected to estimate risk effects . To examine reliability of the pooled risk effects, we additionally performed sensitivity analysis and publication bias tests. Ten studies (1335 HCC patients and 2927 HCC-free controls) were identified for the meta-analysis. We found significantly increased risk of HCC attributable to presence of ADPN I148M polymorphism, with the highest risk associated with the M/M genotype under the recessive model of inheritance . The significant increase persisted in Caucasian and African when data were stratified by ethnicity. Subgroup analysis according to source of controls revealed similar risk effects. Our meta-analysis indicates that I148M polymorphism in the ADPN gene may independently contribute to the progression of HCC irrespective of the etiologies. The same hormonal pathways regulating fat metabolism in the liver function as mediators of ADPN activities.12 A common nonsynonymous SNP (I148M) located at exon 3 of ADPN with a C to G substitution has been consistently associated with liver-related diseases, such as alcoholic liver disease, chronic hepatitis C, and liver fat and fibrosis in nonalcoholic fatty liver disease.15Adiponutrin  or without HCC (n = 275), demonstrating evidence of 2.16 times increased risk in relation to the M/M genotype of I148M polymorphism.Nevertheless, its role in HCC occurrence remains poorly understood. A variety of groups have focused their research on the association of I148M polymorphism with HCC susceptibility, failing to provide convincing evidence for the 17In order to better define the pathological role of I148M polymorphism in HCC occurrence, we decided to perform a meta-analysis of all epidemiological data in agreement with the statement of PRISMA .ADPN, PNPLA3, polymorphism, polymorphisms, and variants. We imposed no limits on language and the minimal sample size. To identify the publications that may be missed during computer-based searches, we manually screened the reference lists of review articles and the original articles with available data on ADPN I148M polymorphism and HCC occurrence. The research was approved by the ethical committee of Shandong Provincial Hospital affiliated to Shandong University. Written consents were obtained from all patients.Literature searches were undertaken through Cochrane Central Register of Controlled Trials, ISI Web of Science, Wiley Online Library, PubMed, and Embase databases to cover all possibly relevant articles. The keywords were comprised of liver diseases, HCC, ADPN I148M polymorphism to HCC risk must be investigated, and no departure from Hardy–Weinberg equilibrium (HWE) in controls, were designed for the studies that could be included in the meta-analysis. We did not consider the studies violating any item listed above.The criteria including the case group consisted of histologically confirmed HCC patients, the controls must be cancer-free subjects, the genetic contribution of If the same patient population was used in several publications by the same authors, we selected the most informative publication containing the largest number of individuals.ADPN I148M polymorphism genotypes. Conflicting evaluations were settled through discussion with the most senior investigator of this study.Based on a standard protocol and data-collection form made according to the inclusion criteria, 2 investigators separately extracted first author’ last name, publication date, total cases and controls, study country, ethnicity, study design (case–control or cohort), source of controls, minor allele frequency (MAF) of controls whenever accessible, source of DNA, mean age, and count of cases and controls with 2 test among control populations to guarantee that all studies included in the meta-analysis were in HWE. Pooled odds ratios (ORs) with 95% confidence intervals (CIs) were computed to estimate the effects of ADPN I148M polymorphism on HCC. The statistical significance was determined with the Z-test. Pooled ORs were obtained by assuming the allele model of inheritance, the homozygous model of inheritance, the heterozygous model of inheritance, the dominant model of inheritance, and the recessive model of inheritance.We first checked the HWE deviation using χ19 was performed to evaluate risk effects of the included studies. Between-study heterogeneity was assessed with χ2-based Q test, and statistical significance level was set at P < 0.05.20 In addition, we used the I2 metric to quantify the variance across studies (heterogeneity),21 and I2 > 50% indicated large heterogeneity. When P < 0.05 or I2 > 50%, we used the random-effects model to calculate the pooled ORs and 95% CIs; otherwise, we selected the fixed-effects model. Subgroup analysis was performed by ethnicity and source of controls.Either the random-effects model or the fixed-effects model derived from the DerSimonian–Laird method and the Mantel–Haenszel method, respectively,22 Stability of the combined risk effects was examined by performing the leave-one-out sensitivity analysis.Evaluation of publication bias was carried out using the funnel plot and Egger's linear regression asymmetry test.P < 0.05 was considered significant.Statistical data were done with Stata v. 12.0 . 31 after excluding 65 studies for various reasons , as shown in Figure We finally identified 10 studies published over the past 4 years in the meta-analysis,ADPN I148M polymorphism using different genotyping assays and 4 groups did not report the DNA sources. The mean age in all studies was above 50 years. The quality assessment of all studies was showed in Table Among the eligible studies, most (90%) collected Caucasian as subjects and only 1 study addressed the association in African. Controls in 6 studies were HCC-free cirrhosis patients, in 3 were healthy subjects, and in 1 including both. Six groups isolated DNA genome from blood samples to detect the genotype of 2 metric .P = 0.358 and P = 0.282, respectively).We constructed the funnel plot for all genetic models, with no one showing obvious asymmetry  had more than 2-fold greater risk to develop the malignancy as compared to the individuals with the wide-type homozygote (I/I) alone or in combination with the heterozygote genotype (I/M). Similar risk effects were observed in Caucasians, because only 1 small study investigated subjects of African ancestry. Unexpectedly, not only studies based on HCC-free cirrhosis patients but also those on the basis of healthy subjects showed a significantly increased risk of HCC in association with ADPN I148M genotypes. Although these results are obtained based on all published data to date, they remain to be verified in subsequent larger investigations, as we cannot rule out the possibility of false positives due to the currently limited sample.A total of 10 studies, providing 1335 cases and 2927 controls, were included in the comprehensive analysis. The derived risk estimates revealed significant genetic contribution of ADPN I148M genotypes on the development of HCC. In this work, Singal et al demonstrated that PNPLA3 is an independent susceptibility factor for HCC among patients with nonalcoholic steatohepatitis or alcohol-related cirrhosis33; the observation derived from the aggregation of 6 studies 34 seems to less precise relative to our study, where 4 additional data sets were identified31 and 1521 more subjects were analyzed, showing obviously higher risk under the dominant model of inheritance (1.40 vs 2.23). This difference seems to imply that ADPN I148M polymorphism is indeed a risk factor for HCC, and only in a study with a sufficient number, can the exact extent of the risk be eventually determined.Our findings are consistent with an earlier meta-analysis examining the effects of HCC has many possible etiologies, such as excessive tobacco use, HBV or HCV infection, heavy alcohol drinking, and liver-related diseases, possibly leading to substantially different susceptibility across the patients with HCC stemming from different carcinogenic agents. Nonetheless, we were unable to consider the etiology of the investigated cancer when examining the association of interest due to data unavailability. This is the first point that needs to be noted in result interpretation.In addition, risk of HCC, like other cancers, is believed to not be determined by a single agent, but affected by both environmental factors and genetic mutations through complex gene–environment and gene–gene interactions. Thus, the modifying influence resulting from confounding variables should be taken into account to derive a more precise risk effect.ADPN I148M polymorphism on HCC occurrence. The statistical data support that HCC incidence was significantly associated with the nonsynonymous polymorphism. Further studies are warranted to verify our findings and detailed examinations according to HCC etiology and ethnicity are expected to identify the specific at-risk populations.In summary, to the best of our knowledge, this is the largest meta-analysis evaluating the impact of


G20210A polymorphism (rs1799963) within the prothrombin gene is associated with a higher circulation level of prothrombin, thus increasing the likelihood of developing myocardial infarction (MI). Opinions differ regarding the correlation between prothrombin G20210A genotype and MI risk, which prompted us to conduct a meta-analysis to determine this association. PubMed, EMBASE, Web of Science and CNKI were searched for pertinent reports. A total of 34 studies involving 14 611 MI cases and 84 358 controls were analyzed in this quantitative analysis. We found a statistically significant association between prothrombin G20210A polymorphism and MI in the allele model , heterozygote model  and dominant model . The association remains significant in Caucasians but not in non-Caucasians. Moreover, prothrombin G20210A polymorphism increases MI risk in an age-related manner. A further significant association was found in a subpopulation younger than 55 years . Sensitivity analysis and publication bias analysis revealed stable and statistically robust results. Our meta-analysis demonstrated that prothrombin G20210A polymorphism may represent a risk factor for MI. Myocardial infarction (MI) is a highly lethal disease in developed countries, usually caused by pathological coronary artery occlusion. However, unstable atherosclerotic plaque is not destined to cause MI. Following plaque rupture, coronary thrombosis involving platelet adhesion, coagulation cascade activation and subsequent thrombus formation plays a fundamental role in the clinical progression to MI. Over-activated procoagulation, acquired from complex patterns of inheritance or environmental risk factors, has more propensity to induce localized coronary occlusion. One of the most concerning etiological factors is single-nucleotide polymorphisms (SNPs) in coagulation and fibrinolytic systems.1. Numerous studies have demonstrated that a G20210A variant (rs1799963) of the prothrombin gene (i.e. a G to A substitution at position 20 210 in the 3′-untranslated region) is associated with elevated circulation levels of prothrombin4 and risk of venous thromboembolism6. Nevertheless, the detrimental role of prothrombin G20210A is controversial in MI8. The reason for this confounding phenomenon may be diversity of the ages of patients recruited in the previous studies. Inherited defects in coagulation may contribute more to MI in the young than in the elderly9. In comparison to MI in the elderly, premature myocardial infarction might be more predisposed to a SNP such as prothrombin G20210A because of relatively less environmental exposure. To draw a conclusion about the clinical impact of the prothrombin G20210A with regard to MI risk within a given age stratum, we gathered data from a number of published studies and performed a meta-analysis.Prothrombin, also called factor II, is the precursor of the thrombin, a key enzyme acting as a procoagulant, through platelet activation and the generation of fibrin and factors Va, VIIIa, and XIII39.As depicted in Fig. 6 and 33 cross-sectional case–control studies. Twenty-two studies included Caucasian patients39, 9 included non-Caucasians32, and 3 did not depict ethnicity of subjects31. As the study population overlapped in two reports39, Maren’s data39 were evaluated only in the >55-year-old subgroup analysis and not in the overall analysis.There was one prospective study27, which consequently was included in the allele model analysis only. Because there was only one eligible study19 of homozygote and recessive models in non-Caucasian subgroup analysis, homozygote and recessive analyses were not performed in the non-Caucasian subgroup.A total of 14 611 cases and 84 358 controls were included in the meta-analysis. Main characteristics, including first author, publication year, study country, ethnicity, age category of MI patients, source of controls, genotyping method, genotype frequencies, and minor allele frequencies are presented in Supplementary Table Overall, a meta-analysis of 13 488 cases and 77 085 controls in 33 studies was performed. The mean age was 48.41 years in cases and 47.16 years in controls. The association between prothrombin G20210A variant adenine (A) allele distribution and MI risk was calculated in comparison with wild-type guanine (G) allele in the allele model (A vs. G). Similarly, the association between specific genotype frequency and MI risk was assessed in the dominant model (AG + AA vs. GG), recessive model (AA vs. AG + GG), homozygote model (AA vs. GG), and heterozygote model (AG vs. GG), respectively. A statistically significant association was found in the allele model  but not in non-Caucasians . Table 28 (on 5 294 patients and 8 149 controls) provided data in subjects younger than 55 years. Four studies39 recruited 2 174 MI patients and 8 261 MI-free controls older than 55 years, respectively. The mean age was 41.75 years in cases and 42.03 in controls in the subgroup ≤55 years old, and 64.33 years in cases and 56.96 in controls in the subgroup >55 years old. Subgroup meta-analysis revealed that a significant association was found in three genetic models between prothrombin G20210A and MI in patients younger than 55 years  but not in individuals older than 55  , 0.62 (homozygote model), 0.13 (heterozygote model), 0.13 (dominant model), and 0.62 (recessive model) in Begg’s funnel plot Fig.  and 0.272. The prothrombin G20210A SNP has been correlated with increased levels of prothrombin in circulation and is clearly associated with an increased risk of venous thrombosis2. Substantive studies over the past decade have been making efforts to unmask the correlation between the G20210A SNP and MI. However, previous conflicting reports failed to draw a consistent conclusion, thereby making the association between the G20210A SNP and MI elusive39. The variance may result from different study designs and distinct ethnic backgrounds of subjects. Furthermore, several studies have reported that the observed association may be affected by other risk factors, such as age at onset. For instance, Butt et al.13 observed that the risk of MI was not elevated in prothrombin G20210A carriers beyond 50 years old, whereas carriers younger than 50 had a 5.6-fold higher risk of MI than age-matched non-carrier control subjects (p = 0.04). Thus, additional risk factors may influence the risk associated with this prothrombin G20210A variation. More importantly, inconsistency in the age of analyzed subjects may also cause discrepancy, as MI events in genetically vulnerable subjects occur much earlier than in normal counterparts40. Two previous systematic reviews41 analyzing the association between G20210A polymorphism and MI risk have drawn similar conclusions, namely that G20210A has no significant association with MI risk in the overall population at all ages, but does have a significant association with MI risk  or risk tendency  in persons younger than 55 years. To analyze whether MI is susceptible to prothrombin G20210A polymorphism, we performed this meta-analysis of 34 published studies to assess the association between prothrombin G20210A polymorphism and MI risk.Myocardial infarction is pathologically caused by coronary thrombotic occlusion following rupture of atherosclerotic plaques. Prothrombin is the precursor of thrombin, which plays a pivotal role in the physiological process of coagulation. Located on chromosome 11, the G20210A variant (rs1799963) substitutes a single base of adenine (A) for guanine (G) at position 20 210 in the 3′-untranslated region of the prothrombin geneIn the current systematic review we extracted allele and genotype frequency of the prothrombin G20210A polymorphism in MI patients and corresponding controls, and estimated the pooled ORs and 95%CI in five genetic models. Our integrated results showed that the G20210A polymorphism was significantly associated with MI risk in the allele model, heterozygote model, and dominant model. The tendency was similarly observed in Caucasians but not in non-Caucasians. The point estimates of the ORs in non-Caucasians, while not significant, are actually higher than those in Caucasians in three models. In subgroup analysis stratified by age, the MI risk in allele, heterozygote and dominant models was even higher in subjects ≤55 years old than in the overall population, but was not statistically significant in those older than 55.41. Boekholdt et al.41 evaluated 4 studies and concluded that G20210A polymorphism was not significantly associated with MI risk . Burzotta et al.7 assessed 13 studies and produced similarly insignificant results . This inconsistency between prior and present outcomes may have at least two possible explanations. Firstly, the two previous meta-analyses enrolled 4 and 13 studies, respectively, whereas 34 studies are included herein. These additional studies with larger numbers of subjects may have enhanced the soundness of the conclusion. Secondly, there may be an age discrepancy between the present and previous meta-analyses. For instance, MI patients younger than 55 years accounted for 16.9% of the total (624/3 687) in Burzotta’s meta-analysis7, whereas in the present study this proportion was 39.2% (5 294/13 488). According to age-related subgroup analysis in previous reviews41 and the present study, prothrombin G20210A polymorphism markedly increases MI risk in younger people. The discrepancy in age may help to explain the disagreeing conclusions.In the present meta-analysis of 34 studies we drew the conclusion that prothrombin G20210A SNP was significantly associated with overall MI risk at all ages, contrasting with results from two previous independent meta-analysis reports7. Premature myocardial infarction (PMI) commonly indicates a first onset of myocardial infarction before the age of 55 years in males and 65 years in females. Considering less exposure to long-term environmental risk factors, MI events in young individuals are more likely attributable to genetic susceptibility than those in elderly counterparts40. The dominant role of acquired risk factors in MI, such as hypercholesterolemia and smoking, attenuates the effects of genetic prothrombin G20210A in the elderly9. Several genetic mutations affecting coagulation proteins have been suggested as PMI risk factors19.An important conclusion from the present research is that G20210A polymorphism correlated with MI in young patients but not in the elderly, consistent with the previous meta-analysis42 of 33 cases of fatal MI and 165 controls showed a noticeable trend toward an association between the GA genotype and fatal MI . (2) Statistically significant association between prothrombin G20210A and MI risk was not observed in the homozygote model  and the recessive model , which seemed to conflict with the findings in the allele model , heterozygote model  and dominant model . The reason for this conflict may be that AA genotype frequency in the overall cases and controls was extremely low (0.07% and 0.04%), which resulted in a population inadequate for researching the association with MI risk. Although two independent researchers searched the databases for published articles and recorded data from eligible studies, it was possible that some articles might not be included in the present study because: (a) the possibility of publication bias could not be eliminated even though no evidence of publication bias was detected through Begg’s funnel plot and Egger’s linear regression method; (b) the language was restricted to English and Chinese. The relatively small sample size and number of included studies for gene-susceptibility investigation might distort the stability of the results regarding homozygote and recessive models. The true detrimental role of homozygote genotype AA in MI may need to be proved in studies with a larger population. (3) Moreover, a potential age discrepancy existed in the overall population analyzed, as there was a greater proportion of young individuals. Subpopulations younger than 55 years were available in 20 studies (containing 5 294 cases and 8 149 controls), whereas those older than 55 were from only 4 studies (containing 2 174 cases and 8 261 controls). The mortality rate of MI in young patients is far less than that in elderly patients, demonstrating the possibility that more elderly patients are lost through fatal MI43.The association between prothrombin G20210A polymorphism and susceptibility to MI has been carefully investigated. Notwithstanding, in addition to the inevitable clinical heterogeneity in most systematic reviews, we must acknowledge some limitations of this meta-analysis. (1) We cannot exclude the potential for a survival bias. Most of the included studies were designed to recruit survivors of MI as a case group. It cannot be excluded that patients who had died during the acute phase of fatal MI more often carried the prothrombin G20210A allele mutation. A necropsy studyIn conclusion, this comprehensive meta-analysis of 34 studies shows evidence supporting a causal association between prothrombin G20210A SNP and vulnerability to myocardial infarction. The risk was higher in the population younger than 55 years, but was not observed in the subjects older than 55. Larger population studies are urgently needed to validate our current findings and address the pathogenic role, which remains poorly understood.44. PubMed, EMBASE, Web of Science and China National Knowledge Infrastructure (CNKI) databases were searched from inception to 30 April, 2016 without language restriction. We used the combination of the search terms:  AND (“polymorphism” OR “polymorphisms” OR “mutation” OR “gene polymorphism” OR “SNP” OR “Single Nucleotide Polymorphism” OR “genotypes” OR “genotype” OR “variants” OR “variant”) AND (”Prothrombin” OR “Factor II”). The searches were complemented by checking the references of the retrieved articles.The PRISMA statement was followed in this meta-analysisEligibility criteria included the following: (1) case-control studies focused on association between prothrombin G20210A and myocardial infarction; (2) at least 30 patients were recruited in the case group; (3) when recruited patients overlapped with those in another study, the study with the largest number of individuals was selected; (4) if a study did not meet the inclusion criteria but a subgroup of the subjects qualified, only the subgroup was included; (5) studies that did not provide sufficient data following contact with authors by e-mail were excluded; (6) case reports and reviews were excluded. There was no notable deviation from Hardy–Weinberg equilibrium (HWE) in the control group.45. Studies that were awarded 5 stars or more, considered as of medium to high quality, were included in the present study. The following information in studies was extracted by two independent researchers: first author, year of publication, ethnicity, country, age of MI patients, source of controls, genotyping method, genotype frequencies, number of cases and controls analyzed, mean value of age.Two independent researchers (J.S. and S.L.) separately assessed the quality of the included studies using the Newcastle–Ottawa Scale (NOS) criteriaR programming language (version 2.13.1) was used to analyze the selected data. The pooled ORs were calculated for the allele model , dominant model (AG + AA vs. GG), recessive model (AA vs. AG + GG), homozygote comparison (AA vs. GG), and heterozygote comparison (AG vs. GG), respectively. After adjusting for multiple testing with Bonferroni’s method, a p value of any comparison less than 0.01 (=0.05/5) was considered statistically significant. When calculating the OR in one genetic model, we excluded studies with no events in either MI or control groups. When only one group contained no events, we added 0.5 to each cell of the 2 × 2 table for the trial before analysis. HWE for each study was determined by Chi-square test.2 statistics and Cochran’s Q statistics for each meta-analysis. I2 more than 50% or a p value less than 0.1 was considered heterogeneous among included studies. In such a case, the REM was used to summarize the values for each study, otherwise the FEM was applied. Begg’s and Egger’s tests were performed to assess publication bias, whereby a p value of less than 0.05 indicates an existence of publication bias among included studies.Tests for study heterogeneity were performed using Higgins’ ISupplementary Information


Overweight and obesity is a major public health concern that includes associations with the development of cardiovascular disease (CVD) risk factors during childhood and adolescence as well as premature mortality in adults. Despite the high prevalence of childhood and adolescent obesity as well as adult CVD, individual studies as well as previous systematic reviews examining the relationship between childhood obesity and adult CVD have yielded conflicting results. The purpose of this study was to use the aggregate data meta-analytic approach to address this gap.z score was calculated for each study for each outcome. Pooled effect sizes were calculated using random-effects models while risk of bias was assessed using the STROBE instrument. In order to try and identify sources of heterogeneity, random-effects meta-regression was also performed.Studies were included if they met the following criteria: (1) longitudinal and cohort studies (including case-cohort), (2) childhood exposure and adult outcomes collected on the same individual over time, (3) childhood obesity, as defined by the original study authors, (4) English-language articles, (5) studies published up to June, 2015, (6) one or more of the following CVD risk factors , (7) outcome(s) not self-reported, and (8) exposure measurements (child’s adiposity) assessed by health professionals, trained investigators, or self-reported. Studies were retrieved by searching three electronic databases as well as citation tracking. Fisher’s r to Of the 4840 citations reviewed, a total of 23 studies were included in the systematic review and 21 in the meta-analysis. The findings suggested that childhood obesity is significantly and positively associated with adult SBP , DBP , and TG , and significantly and inversely associated with adult HDL . For those studies that adjusted for adult body mass index (BMI), associations were reversed, suggesting that adult BMI may be a potential mediator. Nine studies had more than 33% of items that placed them at an increased risk for bias.The results of this study suggest that childhood obesity may be a risk factor for selected adult CVD risk factors. However, a need exists for additional, higher-quality studies that include, but are not limited to, both unadjusted and adjusted measures such as BMI before any definitive conclusions can be reached.CRD42015019763.PROSPERO 2015: The online version of this article (10.1186/s12889-017-4691-z) contains supplementary material, which is available to authorized users. Overweight and obesity during childhood and adolescence is a major public health problem. One of the important health implications of childhood and adolescent obesity includes the development of cardiovascular disease (CVD) risk factors during childhood and adolescence –4. CardiRecent systematic reviews suggest that the relationship between childhood obesity and adult high BP or poor lipid profile is weak, possibly because the results are confounded by adult obesity , 21. In Systematic reviews with meta-analysis occupy the highest levels of evidence in the hierarchy of study designs . This stThis study was conducted and reported according to the Cochrane Collaboration’s recommendations and guidelines for conducting systematic reviews and meta-analyses for observational studies , 33 as wThe a priori inclusion criteria for this meta-analysis included: (1) longitudinal and cohort studies (including case-cohort), (2) childhood exposure and adult outcomes collected on the same individual over time, (3) main exposure variable of the child’s overweight and obesity status , waist-to-hip ratio, visceral adipose tissue, skin fold thickness, body weight, BMI z-score, BMI or other measures used to assess overweight and obesity in populations) , (4) stu“(obesity OR obese OR overweight OR fat OR adipos* OR “body mass index” OR BMI) AND (child* or adolesc*) AND (“blood pressure” OR hypertension OR cholesterol OR lipid OR lipids OR lipoprotein OR lipoproteins OR cardiovascular) AND  AND adult* AND (human OR humans)”.Each search was conducted separately and then downloaded as a separate file using Endnote X7 [Studies were retrieved using three electronic databases: (1) PubMed (MEDLINE), (2) Web of Science, and (3) Scopus. In addition, cross-referencing from the bibliographies of all retrieved articles (citation tracking) was conducted. An information retrieval specialist  assisted in the planning of the literature search and in identifying and creating correct Boolean operators and search strings for the different electronic database searches . The Pubdnote X7 . The firIn order to minimize selection bias, two researchers (AU and CL) independently screened studies for eligibility by reviewing the titles and abstracts of articles based on the pre-defined eligibility criteria. If the inclusion or exclusion criteria could not be decided based on the title and abstract, full-text articles were retrieved and the decision was made accordingly. After independent study selection was performed, the two reviewers met and reviewed every selection for agreement and discrepancies were resolved by consensus. Using Cohen’s kappa statistic, the overall agreement rate prior to correcting discrepancies was 0.75 . If a dePrior to data abstraction, a detailed codebook that could hold up to 200 items per study was developed by the research team in Microsoft Excel (version 2011) . The codRisk of bias was assessed using the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) instrument . The STRWe conducted an aggregate data meta-analysis for all seven outcomes: (1) SBP, (2) DBP, (3) TC, (4) HDL, (5) LDL, (6) non-HDL and (7) TG. Seven separate Microsoft Excel sheets were generated for all seven outcomes. Any outcome with only one study was excluded from the meta-analysis. Each outcome was further analyzed separately if it was adjusted for adult adiposity.The primary outcome for this study was the correlation between childhood adiposity and adult CVD risk factors. The correlation coefficient ‘r’ was transformed using Fishers r to z transformation. The a priori plan was to use risk ratio (RR) as our ES in order to examine the association between childhood obesity and selected adult CVD risk factors. However, because most studies reported a correlation between two continuous variables, a post hoc decision was made to use correlation statistics (Fishers r to z score) instead of RR to serve as the main ES index . All othStandardized beta coefficients from individual studies were used as correlation statistics based on previous research showing that the correlation between a beta coefficient and correlation coefficient is linear, having a correlation of 0.84 if the coefficients reside in the interval ± 0.50 . StudiesResults for the association between childhood obesity and selected CVD risk factors  were pooled separately using random-effects, method-of-moments models . The corI2 [I2 was classified as trivial (0%–25%), low (25.1%–50%), moderate (50.1%–75%), or high (75.1%–100%) [Heterogeneity was examined using the Q statistic while inconsistency was assessed using I2 , 57. StaSmall-study effects (e.g. publication bias) was examined qualitatively using the funnel plot and quantitatively using Egger’s linear regression test [. If the intercept was not significantly different from zero, it was assumed that there was no evidence of funnel plot asymmetry [. Influence analysis with each study deleted from the model once was conducted in order to examine the effects of each study on the overall results [umulative meta-analysis, ranked by the year the study started, was used to examine the accumulation of findings over time [sensitivity analysis was also performed by pooling the ESs from studies that only used childhood BMI as the exposure. This was performed in order to determine if any differences existed in the pooled results that included any definition for childhood exposure, including BMI.ion test . As recoion test . The Eggsymmetry , 60, 61. results . Cumulatver time . A sensiBecause of missing data for different predictor variables from different studies, simple weighted least squares meta-regression  was used to examine the relationship between each outcome and selected covariates. Meta-regression is analogous to individual study regression except that the outcome variable is the effect estimate, i.e., unit of analysis is the study, rather than individual participant scores . The sloA general description of the characteristics of each study is shown in Table The length of follow-up for the studies ranged from 4.5 to 60 years. As most of the studies were prospective longitudinal studies, the number of subjects at baseline was often greater than the number of participants at follow-up due to loss at follow-up. Seven studies included information on loss to follow-up , 77, 82.With respect to exposure variable measurements, none were self-reported. Most studies used BMI as a measure of adiposity in addition to other measures used , 81–85. Sixteen studies examined the association between childhood obesity and adult SBP , 82, 85,As previously stated, a description of the participant characteristics is shown in Table The participants’ ages at baseline when the exposure was measured ranged from 2 to 18 years. The age at follow-up when the outcome was measured also varied substantially between studies, ranging from 19 to 62 years. Some studies used one childhood age or an average of a range of childhood ages as their exposure, while other studies categorized children based on different age groups. Furthermore, some studies used longitudinal data for the same children over time.Most studies provided some level of information on comorbid conditions as adults. These included hypertension , arterioOverall study-level risk of bias results are shown in Fig. The results of the random-effects meta-analysis for the association between childhood obesity and adult CVD risk factors are presented in Tables p = 0.42). With each study deleted from the model once, results remained statistically significant across all deletions, ranging from 0.10 to 0.11. Cumulative meta-analysis, ranked by the year the study started, demonstrated that results have been statistically significant since examining the birth cohort of 1958. Random-effects meta-regression revealed statistically significant evidence for an association between pooled ES (Fisher’s r to z score for the association between childhood obesity and adult SBP) and baseline age  only.Overall, there was a statistically significant and positive association between childhood adiposity and adult SBP Table . Heterogp = 0.002; I2 reduced from 89% to 69%) as well as length of follow-up (follow-up age – baseline age) .When examining studies that adjusted for adult BMI, a statistically significant and negative association was observed between childhood adiposity and adult SBP Table . Heterogp = 0.37). With each study deleted from the model once, results remained statistically significant across all deletions. The associations ranged from 0.09 to 0.11. Cumulative meta-analysis, ranked by the year the study started, demonstrated that results have been significant since examining the birth cohort of 1966. Random-effects meta-regression revealed statistically significant evidence for an association between pooled ES (Fisher’s r to z score) for the association between childhood obesity and adult DBP and baseline age .Overall, there was a statistically significant and positive association between childhood adiposity and adult DBP Table . Heterogp < 0.0001; I2 reduced 92% to 58%), length of follow-up  as well as year of study onset .When examining studies that adjusted for adult BMI, a statistically significant and negative association was observed between childhood adiposity and adult DBP Table . Heterogp = 0.04; I2 reduced from 91% to 76%), length of follow-up , as well as the year of study onset .Overall, there was a positive association between childhood adiposity and adult TC Table . Howeverp < 0.0001; I2 reduced from 86% to 0%) and sex .When examining studies that adjusted for adult BMI, a negative and non-significant association was observed between childhood adiposity and adult TC Table . The 95%p = 0.0001; I2 reduced from 94% to 39%), length of follow-up , and the year of study onset .Overall, there was a positive association between childhood adiposity and adult LDL Table . HoweverI2. In addition, 95% PIs were non-significant. Because there were less than 10 studies, small-study effects were not assessed. With each study deleted from the model once, results remained statistically significant across all deletions. The associations ranged from approximately −0.07 to −0.09. Cumulative meta-analysis, ranked by the year the studies started, demonstrated that results have been statistically significant since the first study was conducted in 1947. Random-effects meta-regression revealed no statistically significant evidence for an association between pooled ES (Fisher’s r to z score for the association between childhood obesity and adult LDL for studies that adjusted for adult BMI) and sex. Because of insufficient data, none of the other variables were examined.When examining the three studies that adjusted for adult BMI, a statistically significant and negative association was observed between childhood adiposity and adult LDL Table . HeterogOverall, there was a statistically significant and negative association between childhood adiposity and adult HDL Table . Heterogp = 0.004; I2 reduced from 94% to 74%), length of follow-up , sex , and risk of bias assessment for low risk .When examining studies that adjusted for adult BMI, a positive but non-significant association was observed between childhood adiposity and adult HDL Table . Heterogp < 0.0001; I2 reduced from 83% to 5%), length of follow-up , and risk of bias assessment for low risk .Overall, there was a statistically significant and positive association between childhood adiposity and adult TG Table . Heterogp = 0.05; I2 reduced from 95% to 92%), length of follow-up  and sex .When examining studies that adjusted for adult BMI, a negative association was observed between childhood adiposity and adult TG Table . HoweverInsufficient data were available to analyze non-HDL.The results of our sensitivity analysis for those outcomes using only BMI as the exposure showed similar findings Table . The finThe purpose of this study was to conduct a systematic review and meta-analysis of studies that have examined the association between childhood obesity and adult CVD risk factors . The overall unadjusted findings suggest that childhood obesity is significantly and positively associated with adult SBP, DBP and TG and significantly and negatively associated with adult HDL. This interpretation is supported by: (1) 95% CI for overall results that do not include the null, (2) consistency of overall results when each study was deleted from the model once , (3) significance of results over a long time period in which the included studies were conducted , and (4) non-significant small study effects.When examining studies that adjusted for adult obesity, the overall findings suggest that the association was significant and negative for SBP, DBP, and LDL while the associations between childhood obesity and adult HDL and TG became non-significant when adult BMI was accounted for. However, it is important to point out that less than one third of studies adjusted for adult adiposity measures , 78, 85.Several factors need to be considered when examining the results of this study. First, we used random-effects models that incorporate heterogeneity into the analysis. However, based on the fixed-effect model, we observed a moderate to large amount of heterogeneity in the vast majority of outcomes assessed. While a random-effects model incorporates heterogeneity into the analysis, it does not explain the sources of heterogeneity. Second, the 95% PIs were not statistically significant as they overlapped the null (0) and thus, give us less confidence in the overall results of the study based on 95% PI’s , 87. ThiMore specifically, results from our meta-regression analyses revealed that the association between childhood obesity and adult SBP and DBP increases as the baseline age increases. For TC, HDL, and LDL the association decreases as the follow-up age and length of follow-up increases and for TC and LDL the association increases as the year of study onset increases. For TG, only the association increased as the percent of low risk of bias increased. For studies that adjusted for adult BMI, the association between childhood obesity and adult TC increased as baseline age increased. For SBP, DBP, and HDL the association decreased as the follow-up age and length of follow-up increased and for HDL the association increased as the year of study onset increased. The association was lower in males compared to females for TC and HDL, and higher in males compared to females for TGs. For HDL, only the association increased as the percent of low risk of bias increased. However, one unusual finding was the association between childhood obesity and adult TG (adjusted for adult BMI) that increased with the increase in the follow-up age/length of follow-up. We hypothesize that this odd finding may be due to the play of chance given all the tests that were conducted. However, the results of the meta-regression tests should be interpreted with caution since they are considered observational inquiries to generate hypotheses about potential sources of heterogeneity  i.e. to As previously discussed, four systematic reviews , 21–23 pP = 0.009) [A quantitative analysis by Juonala et al.,  used dat= 0.009) . For dys= 0.009) . Some ofThe result of the current systematic review with meta-analysis has several implications for reporting and conducting future longitudinal studies. First, based on the STROBE instrument, it is recommended that future longitudinal studies improve their reporting with respect to several potential sources of bias. These include: 1) describing any efforts to address potential sources of bias, (2) explaining how missing data were addressed, (3) explaining how loss to follow-up was addressed, 4) describing any sensitivity analyses conducted, (4) reporting the numbers of individuals at each stage of the study, (5) providing reasons for non-participation at each stage, (6) considering the use of a flow diagram, (7) describing the characteristics of study participants, (8) considering translating estimates of relative risk into absolute risk for a meaningful time period, and (9) reporting any other analyses conducted . Because longitudinal studies have a criterion for initially selecting participants that choose to participate or not, have varied response rates, different numbers of participants at baseline and follow-up, as well as varied participation and response rates at follow-up time point(s), it is important to provide this information using a flow diagram. However, only one study used a flow diagram. Therefore, it is suggested that future longitudinal studies include a flow diagram to clearly demonstrate their study design, participation and response rates. Second, complete information on the population characteristics should be presented, usually in Table  describi describiThe result of the current systematic review with meta-analysis has relevant implications for practice. Overall, it appears that childhood obesity is positively associated with adult SBP, DBP, and TG and negatively associated with adult HDL. Although we did not evaluate the likelihood of a causal association using Hill’s criteria, several of these criteria  suggest childhood obesity as a plausible risk factor for adult CVD risk factors . Given tTo the best of our knowledge, this is the most recent and complete study that has systematically appraised studies examining the associations included in this systematic review with meta-analysis. It is based on a greater number of studies (published up to June 2015) that included both crude associations as well as studies that adjusted for adult adiposity. This work also included any definition of adiposity and measure that was utilized for the exposure. This may be particularly relevant since it has been suggested that BMI is not an ideal marker for adiposity , 28. FroAlthough we performed the main meta-analysis using studies that utilized varied childhood adiposity measures, the results of our sensitivity analysis using only BMI as the exposure showed similar findings Table . In addiThe results of the current meta-analysis should be viewed with respect to the following potential limitations: (1) only one third of the included studies adjusted for adult BMI, (2) some of the pre-planned analyses to identify sources of heterogeneity were not performed due to lack of data, (3) the sample sizes for some of the analyses may have been underpowered to find a true effect, suggesting that future original studies may want to include such information, (4) due to the small sample sizes for some analyses, small-study effects  were not conducted, (5) meta-analysis inherits the limitations of the original studies included, (6) a lack of empirical evidence, including assessment tools, for assessing study quality, especially given the difficulty in differentiating between quality of reporting and quality in the conduct of a study , 47–51 aThe results of this study suggest that childhood obesity may be a risk factor for selected adult cardiovascular disease risk factors. However, a need exists for additional, well-designed studies that include both unadjusted and adjusted data before any definitive conclusions can be reached.Additional file 1:Search strategy to find existing systematic reviews with or without meta-analysis in PubMed (MEDLINE). (DOCX 173 kb)Additional file 2:This file provides the previous systematic reviews and meta-analysis examining the association between childhood obesity and selected adult CVD risk factors including AMSTAR assessment of the previous systematic reviews and meta-analysis. (DOC 131 kb)Additional file 3:PRISMA checklist (DOC 69 kb)Additional file 4:Search strategy for electronic database. This file provides the search strategy used for the (1) PubMed (MEDLINE), (2) Web of Science, and (3) Scopus. (DOCX 574 kb)Additional file 5:Excluded studies, including reasons. This file provides a reference list of all excluded studies, including the reasons for exclusion. (XLSX 900 kb)Additional file 6:STROBE Checklist (DOCX 37 kb)Additional file 7:Influence meta-analysis, cumulative meta-analysis and funnel plots (DOCX 3535 kb)Additional file 8:Meta-Regression results for the association between childhood adiposity and adult CVD risk factors (DOCX 47 kb)


Adequate sleep is important for good health, but it is not always easy to achieve because of social factors. Daytime napping is widely prevalent around the world. We performed a meta-analysis to investigate the association between napping (or excessive daytime sleepiness: EDS) and the risk of type 2 diabetes or metabolic syndrome, and to quantify the potential dose-response relation using cubic spline models. Electronic databases were searched for articles published up to 2016, with 288,883 Asian and Western subjects. Pooled analysis revealed that a long nap (≥60 min/day) and EDS were each significantly associated with an increased risk of type 2 diabetes versus no nap or no EDS  for a long nap and 2.00 (1.58–2.53) for EDS). In contrast, a short nap (<60 min/day) was not associated with diabetes (p = 0.75). Dose-response meta-analysis showed a J-curve relation between nap time and the risk of diabetes or metabolic syndrome, with no effect of napping up to about 40 minutes/day, followed by a sharp increase in risk at longer nap times. In summary, longer napping is associated with an increased risk of metabolic disease. Further studies are needed to confirm the benefit of a short nap. Adequate sleep is important for maintaining good health, along with a balanced diet and exercise. Several recent studies have shown that the relation between the duration of nocturnal sleep and the risk of type 2 diabetes15A nap is defined as a short sleep that is typically taken during daylight hours, and the habit of napping is prevalent worldwide. While a daytime nap is usually brief, it can range from a few minutes to several hours. In addition, the frequency varies from occasional naps to several times daily for habitual nappers. Some people take a nap because of excessive daytime drowsiness resulting from a sleep disorderA short nap (<30 min) promotes alertness, reduces sleep deficits, and enhances performance and learningWe recently published a dose-response meta-analysis that identified a J-curve relationship between nap time and cardiovascular disease based on 1.5 million person-years of dataRecent epidemiological studies on the relation between daytime napping and diabetes or metabolic syndrome have yielded conflicting results1516171819212223We performed a literature search (up to 31 January 2016) of the Medline, Cochrane Library, and Web of Science databases to identify observational studies examining the association between napping and/or excessive daytime sleepiness and the risk of type 2 diabetes and/or metabolic syndrome. The details of the search terms are shown in We selected studies that reported risk estimates for type 2 diabetes and metabolic syndrome in relation to daytime napping and excessive daytime sleepiness in the general population, and that provided point estimates of odds ratio with the 95% confidence interval or standard error for qualitative assessment. We performed a combined meta-analysis of the relation between napping and diabetes or excessive daytime sleepiness and diabetes, because napping and daytime sleepiness have overlapping features and are often handled in the same way clinically by physicians and co-medical staff. We defined the presence of daytime napping on the basis of an affirmative answer to questions such as “Do you take a daytime nap?” or “Do you sleep during the day?” Excessive daytime sleepiness was identified by affirmative answers to questions like “Do you have a problem with sleepiness during the daytime?” We excluded studies on type 1 diabetes. The studies that we selected followed the relevant local rules for ethics and data protection. Studies that did not report risks independently stratified by at least 3 nap time categories  were excluded from this meta-analysis.We extracted information on the characteristics of each study , the subject characteristics , the extent of exposure to napping , the validity of the method used for assessment of napping (and excessive daytime sleepiness), the validity of the method used for assessment of the outcome (diabetes and metabolic syndrome), and the validity of the analytical methods . Some studies reported data separately for men and women, so we treated each of these cohorts as an independent report and extracted data separately. If more than one study covered the same cohort, only the report containing the most comprehensive information was used to avoid analysis of overlapping populations.To ascertain the validity of the studies, the quality of each report was appraised with reference to the STROBE statementet al.χ2 test and the I2 test were used to evaluate heterogeneity among the studiesThe pooled odds ratio (OR) and its 95% confidence interval (CI) was employed to assess associations in all of the studies, except for use of the prevalence ratio by Stang et al.et al.We included the odds ratios of the longest (or shortest) groups in the pairwise meta-analysis  if the published studies reported odds ratios for various nap times . Stratified analyses were also performed with stratification by study location, study score, and study type, and we used the method of Altman et al.Possible publication bias was evaluated by creating a funnel plot of the effect size for each study versus the standard error. Then asymmetry of the funnel plots was assessed by performing Begg’s testet al.The midpoint of the upper and lower borders was set as the median dose for each category if the median or mean exposure per category was not reported. If the highest category was open-ended, the midpoint of the category was set at 1.25 times the lower border. For the lowest nap category, we set the median at 0.5 times the cut-off point . In the study of Lam Because the estimators in the random effects cubic spline model were found to be unstable due to lack of power, we used a fixed effects model to evaluate the dose-response relation. All statistical analyses were performed with Stata V.14.0 software . P values of less than 0.05 were considered significant.This research was carried out according to a predetermined protocolet al.151617181920212223151920et al.17181921p < 0.01, I2 = 89%]) compared with the absence of these factors, while a shorter nap time (<60 min/day) was not associated with an increased risk of diabetes . Similar results were obtained when analyses were performed with stratification by study location, study quality, and study type  (long nap vs. no nap). The funnel plot, Begg’s test, and Egger’s test did not suggest any evidence of publication bias  (udy type . Heterogion bias . In addi = 50%]) .P for non-linearity = 0.003)  . That isp < 0.01, I2 = 10%]), while a shorter nap time (<60 min/day) was not associated with an increased risk of metabolic syndrome   was associated with a significantly higher risk of metabolic syndrome  . Similar2 = 0%]) . NappingThis meta-analysis showed that either a long nap or EDS was associated with an increased risk of type 2 diabetes and metabolic syndrome compared with not napping or no EDS. In contrast, a short nap was not associated with diabetes or metabolic syndrome. We also demonstrated that nap time may show a J-curve relationship with these metabolic diseases, as has been reported for cardiovascular diseaseAs we have stated2439Moreover, a shorter nap might not have the same negative effects as a longer nap. Several studies have demonstrated that a daytime nap of less than 30 min promotes wakefulness and alertness, reduces sleep deficits, and enhances performance and learning4142434411Intermittent hypoxemia and sleep fragmentation increase sympathetic activity4950et al. reported that inactive people have a higher prevalence of napping compared with active peopleOur present meta-analysis had some limitations. First and importantly, while we analysed results from adjusted models because the original studies were observational investigations, this meta-analysis was mainly based on case-control studies and thus cannot prove causal effects. Moreover, the findings could have been influenced by residual confounders or other biases . Thus, the associations detected in the individual studies and our meta-analysis may have been biased in either direction. Second, a high level of heterogeneity was detected. To explore the reasons for this, we performed analyses stratified by study location, study quality, and study type. However, heterogeneity was not dramatically reduced by these stratified analyses. Because we identified a nonlinear J-curve relation between nap time and diabetes or metabolic syndrome in our dose-response meta-analysis, it is possible that differences of nap time among the studies analyzed might have led to high heterogeneity. Third, there was limited information about the possible contribution of physical activity to diabetes. Leng Ideally, our results should be confirmed by prospective clinical trials, but it seems difficult to perform an interventional longitudinal study of napping in the real world (participants would often be unable to nap for the specified time). Accordingly, we cannot rule out the possibility of “post hoc ergo propter hoc”.Fourth, we might have overlooked some relevant articles, which could have led to selection bias. However, we investigated all of the references cited in each study we selected to avoid this risk as far as possible. Also, our analyses of publication bias did not suggest that unpublished results had been missed. Fifth, all of the studies used structured interviews or self-completed questionnaires to assess nap times, so measurement error is inevitable. However, measurement error with regard to assessment of exposure is more likely to cause attenuation of a true association than to exaggerate a weak association. Sixth, although our funnel plot analyses did not show significant publication bias, the limited number of studies may have diminished the statistical power for detecting heterogeneity. Finally, we employed different measures (“standardized increment of nap time” or midpoint of nap time) when modeling the associations of the dose-response relation in our meta-analysis. However, none of the studies gave precise data on nap times, which raises concern regarding the precision of our analyses. Despite these limitations, we consider that our meta-analysis provides useful evidence regarding the potential influence of napping on metabolic disease.In conclusion, the relation between nap time and the risk of diabetes or metabolic syndrome seems to be described by a J-curve, with a sharp increase in the risk of diabetes or metabolic syndrome at longer nap times. In contrast, a short nap was not associated with an increased risk of either diabetes or metabolic syndrome.Additional research, including large-scale pooling projects, with accurate and detailed measures of napping will be needed to confirm our conclusions and to determine whether adding a nap time score to a conventional risk model improves estimation of the long-term risk of diabetes in the general population, as well as whether a short nap decreases the risk of developing type 2 diabetes.How to cite this article: Yamada, T. et al. J-curve relation between daytime nap duration and type 2 diabetes or metabolic syndrome: A dose-response meta-analysis. Sci. Rep.6, 38075; doi: 10.1038/srep38075 (2016).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


Since a genome-wide association study revealed that Interleukin-23 receptor (IL-23R) gene is a candidate gene for Ulcerative Colitis (UC), many studies have investigated the association between the IL-23R polymorphisms and UC. However, the results were controversial. The aim of the study was to determine whether the IL-23R polymorphisms confer susceptibility to UC.A systematic literature search was carried out to identify all potentially relevant studies. Pooled odds ratios (ORs) with 95% confidence intervals (CIs) were used to estimate the strength of association.P < 0.001). Similarly, meta-analyses of the rs7517847, rs1004819, rs10889677, rs2201841, rs11209032, rs1495965, rs1343151 and rs11465804 polymorphisms also indicated significant association with all UC . Stratification by ethnicity revealed that the rs11209026, rs7517847, rs10889677, rs2201841 andrs11465804 polymorphisms were associated with UC in the Caucasian group, but not in Asians, while the rs1004819 and rs11209032 polymorphisms were found to be related to UC for both Caucasian and Asian groups. However, subgroup analysis failed to unveil any association between the rs1495965 and rs1343151 polymorphisms and UC in Caucasians or Asians.A total of 33 studies in 32 articles, including 10,527 UC cases and 15,142 healthy controls, were finally involved in the meta-analysis. Overall, a significant association was found between all UC cases and the rs11209026A allele (OR = 0.665, 95% CI = 0.604~0.733, The meta-analysis suggests significant association between IL-23R polymorphisms and UC, especially in Caucasians. Ulcerative Colitis (UC) is a chronic relapsing and remitting intestinal inflammatory disorder that is confined to the mucosal and submucosal layers of rectum and colon, invading epithelial lining of the gut , 2. The et al. found that IL-23R was expressed at a significantly elevated level in the peripheral blood of ankylosing spondylitis (AS) patients, and the increased IL-23R levels were driven by large increases in the proportion of γ/δ T cells expressing IL-23R [et al. firstly revealed that IL-23R was a potential candidate gene for UC susceptibility [Interleukin-23 (IL-23) is a pro-inflammation cytokine secreted by activated macrophages and dendritic cells, which can influence the differentiation of native CD4+ T cells into T helper 17 subsets . As a nog IL-23R . IL-17-sg IL-23R . An IL-2g IL-23R . IL-23/Ig IL-23R , 17. Furg IL-23R . A genomtibility . IL-23R tibility . In the tibility –51. Howeet al. have performed a meta-analysis to assess the association between the IL-23R gene polymorphisms and UC risk, but this meta-analysis only included sixteen studies [Meta-analysis is a statistical method of augmenting the effective sample size through merging data from single association studies, thus enhancing the statistical power of the analysis for the estimation of genetic effects . Liu et  studies , and morAccording to the inclusion and exclusion criteria, 32 published articles were included in the meta-analysis. Of the 32 articles, one included two cohorts , thus, e2 = 14.14, I2 = 64.9%, P = 0.015; χ2 = 14.14, I2 = 71.7%, P = 0.007, respectively). For the IL-23R rs1343151 polymorphism, stratification analysis by ethnicity detected significant heterogeneity in the Caucasian group . Sensitivity analysis was adopted to assess the stability of the results and explore potential sources of heterogeneity across studies. We used leave-one-out method by sequentially omitting each study to assess the influence of individual data on the obtained results.  . However, ethnicity-specific analysis failed to identify any associations between the two polymorphisms and UC risk in neither Caucasian nor Asian subjects.Meta-analysis showed significant association between the minor alleles of the rs1495965 and rs1343151 polymorphisms and the risk of UC in the overall population . However, there is no such study performed in the Asian group.No association was found between UC and the IL-23R rs10489629 polymorphism by meta-analyses  and UC. Overall, IL-23R gene polymorphisms (except for rs10489629) are associated with UC risk. We also undertook subgroup analyses by ethnicity for these polymorphisms. Our findings provide strong evidence of a significant association between the IL-23R polymorphisms, with the exception of the rs1495965 and rs1343151 polymorphisms, and UC in Caucasians. The present study showed that the IL-23R rs1495965 and rs1343151 polymorphisms were not associated with UC in Caucasians and Asians. The reasons for this disagreement might arise from the followings. We found that the ORs (95% CIs) of the individual studies , 26 inclThe diverse roles of the same gene polymorphism in subgroup analysis by ethnicity could be ascribed to the followings. It is well known that UC is the most prevalent in North America and Europe, and the least prevalent among African Americans and Asians . There iet al. found the protective allele A of rs11209026 SNP could increase the expression of the soluble form of IL-23R mRNA and diminish the induction of Th17 cells upon IL-23 stimulation [With regard to the functional effects of IL-23R polymorphisms, rs10889677, a variant located in the 3′-untranslated region of the IL-23R gene, have been demonstrated to induce the loss of microRNA regulation and enhance the IL-23R mRNA levels and protein production. In combination with this variant (rs10889677), miRNA-mediated dysregulation of IL-23 signaling strongly associated with UC susceptibility . Furthermulation .et al. has examined the association of the IL-23R gene polymorphisms as a genetic predictor of infliximab responsiveness in UC patients and showed that UC patients carrying the IBD risk-increasing IL-23R polymorphisms were more likely to respond to infliximab than patients carrying the IBD risk-decreasing IL-23R polymorphisms , which suggests that IL-23R gene is related to treatment of UC patient [The strength of our meta-analysis could be summarized as the followings. We did our best to identify as many published articles as we could through various searching approaches. Compared with the previous meta-analysis , the cur patient . TherefoCertainly, some limitations of the present study should be addressed. First, this study could not analyze the potential gene-environment interactions and gene susceptibility haplotypes due to insufficient data. Second, our literature search was only limited on English and Chinese, language bias might exist. Third, significant between-study heterogeneity and publication bias was found in some analyses, which might have an influence on the current study. Fourth, some results are still controversial owing to relatively small sample size in Asians, and the results of meta-analysis were applicable only to the Asian and Caucasian groups.In summary, the current study provides a comprehensive examination of the available evidence for the association between polymorphisms in the IL-23R gene and UC. The meta-analysis suggests that IL-23R gene polymorphisms are associated with UC susceptibility, especially in Caucasians. However, larger sample size studies taking environmental risk factor into account and including other ethnic groups are need to confirm the results from our meta-analysis in the future.A systematic literature search in PubMed database, Elsevier Science Direct, China National Knowledge Infrastructure database (CNKI), Wanfang Database, and Chinese Biomedical database (CBM) was carried out to identify potential articles and references in the identified articles were checked manually to find additional studies concerning the association between polymorphisms in the IL-23R gene and UC risk. The text words were as follows: “ulcerative colitis or inflammatory bowel disease or UC or IBD” and “Interleukin-23 receptor or IL-23R or IL23R” combined with “gene or polymorphism or variant”. The languages were limited to English and Chinese. The last search was updated on May 1, 2016.A study was included in the analysis if a) it was a case-control or cohort study; b) it included original data (independence among studies); c) the studies tested the association between the IL-23R gene polymorphisms  and UC risk, and d) the studies provided sufficient data to calculate the statistical indicators. Studies were excluded if one of the followings existed: a) studies contained overlapping data; b) studies in which family members had been studied because of the analysis based on linkage considerations.P-value.The following data were extracted independently by two authors from each eligible article: first author's last name, publication year, population, ethnicity, numbers of cases and controls, genotyping methods, polymorphisms in the IL-23R gene studied and Hardy-Weinberg equilibrium (HWE) 2 test.Allele frequencies at the IL-23R gene polymorphisms from each single study were determined by the counting method. The pooled odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to assess the strength of association between these gene polymorphisms and UC susceptibility. HWE was examined using the χ2-test based on the Q statistic was applied to examine the heterogeneity of between-study [2I was more than 50% and P value was less than 0.1, the heterogeneity was considered to be significant, and then the random effects model was selected for meta-analysis. Otherwise, a fixed effects model was adopted.The χen-study . If I2 wP value was less than 0.05, statistically significant publication bias might exist [Publication bias was estimated by Egger's linear regression test and the Funnel plot. If the ht exist .P value < 0.05 was regarded as statistically significant.All the statistical analyses of meta-analysis were performed by STATA statistical software . A two-sided