Background:LRRK2 variants have been demonstrated to have distinct distributions in different populations. However, researchers have thus far chosen to focus on relatively few variants, such as R1628P, G2019S, and G2385R. We therefore investigated the relationship between common LRRK2 variants and PD risk in various populations.Methods: Using a set of strict inclusion criteria, six databases were searched, resulting in the selection of 94 articles covering 49,299 cases and 47,319 controls for final pooled analysis and frequency analysis. Subgroup analysis were done for Africans, European/West Asians, Hispanics, East Asians, and mixed populations. Statistical analysis was carried out using the Mantel-Haenszel approach to determine the relationship between common LRRK2 variants and PD risk, with the significance level set at p < 0.05.Results: In the absence of obvious heterogeneities and publication biases among the included studies, we concluded that A419V, R1441C/G/H, R1628P, G2019S, and G2385R were associated with increased PD risk , while R1398H was associated with decreased risk (p: < 0.00001). In East Asian populations, A419V, R1628P, and G2385R increased risk , while R1398H had the opposite effect (p: 0.0005). G2019S increased PD risk in both European/West Asian and mixed populations , while R1441C/G/H increased risk in European/West Asian populations only (p: 0.0004).Conclusions: We demonstrated that LRRK2 variant distribution is different among various populations, which should inform decisions regarding the development of future genetic screening strategies. Parkinson's disease (PD) is one of the most common neurodegenerative diseases, affecting ~2% of people over the age of 60 years, and the most common cause of movement disorders, including bradykinesia, resting tremor, rigidity, and postural instability or gait difficulty. Non-motor symptoms, such as depression, olfactory dysfunction, and constipation are also common in PD  gene, which encodes for a protein containing armadillo (ARM), Ras of complex proteins (ROC), C-terminal of ROC (COR), mitogen-activated protein kinase kinase kinase (MAPKKK), and WD40 domains, in addition to others  and at least four original articles reported on the same variant, and “study types” consisted of original case-control studies, cohort studies  approach was used to define inclusion criteria. For this study, “participants” were patients diagnosed with PD according to accepted standards, such as the UK PD Society Brain Bank Clinical Diagnostic Criteria  were extracted from all selected original studies, and are detailed in the P > 0.1 and I2 statistic ≤ 50%, a fixed-effect model was used, otherwise, a random-effect model was applied for pooled analysis instead. All pooled results were graphed using forest plots and publication biases were showed using funnel plots. Subgroup analysis using the Mantel-Haenszel statistical method was performed to determine how the common LRRK2 variants affect PD risk and the level of significance was set at p < 0.05. Sensitivity analysis was performed by sequentially deleting each included article, and observing how the pooled OR and 95%CI was affected by their removal. Genotype frequency (GF) and minor allele frequency (MAF) were calculated of each LRRK2 variants in our analyses.Statistical analysis was performed using Revman 5.3 software. The pooled analyses were conducted if there were at least four original studies. Meta-analyses were conducted on total populations and subgroup analyses by ethnicity  , while R1398H was associated with decreased PD risk , while R1398H had the opposite effect (p: 0.0005). The G2019S variant was found to increase PD risk in European/West Asian and mixed populations (p: < 0.00001 and < 0.00001), while R1441C/G/H increased risk for European/West Asians only  were found to increase PD risk. Although previous studies have been quite successful at identifying high-frequency risk variants, our findings serve to highlight the fact that even those that occur at lower frequencies, such as A419V, should not be neglected, particularly in East Asian populations, even if clinically significant data may only be accessed in these cases through the use of larger sample sizes that are possible only with collaborative multi-center projects. We also determined that G2019S can increase PD risk in European/West Asian and mixed populations, while R1441C/G/H increases PD risk in European/West Asians only. The pooled analysis and frequency analysis of variants in LRRK2 gene has always posed a challenge for researchers interested in screening for the gene due to its large size. For the sake of improving the efficiency and economy of genetic diagnosis, it is thus of great importance to prioritize the identification of specific variations instead of sequencing the entire gene  scores, high University of Pennsylvania Smell Identification Test (UPSIT) scores and responded well to Levodopa. G2385R carriers tended to have family history, lower Hoehn and Yahr rating (H-Y) and higher Mini Mental State Examination (MMSE) scores. However, both G2019S and G2385R carriers were more likely to develop motor complications than non-carriers (Shu et al., LRRK2 variant to screen for. Additionally, identifying the specific clinical features associated with carrier of particular LRRK2 variants could also be useful for neurologists in prescribing the correct symptomatic treatments.Certain There are, of course, limitations to this study. Firstly, although we have endeavored to include every published case-control study in the meta-analysis, certain pieces of unpublished data, or articles that were written in languages other than English or Chinese may have been omitted unintentionally. Secondly, due to the lack of sufficient data, variants in populations other than East Asians and European/West Asians cannot be analyzed, and it is possible that biases may exist in our pooled analysis of different ethnic groups and in their demographic information, such as age and gender. Thirdly, stratified analysis can inadvertently increase the possibility of there being false positives, especially as the sample size is limited.LRRK2 variants A419V, G2019S, R1441C/G/H, G2385R, and R1628P were associated with increased PD risk while R1398H was associated with decreased risk. In East Asian populations, A419V, G2385R, and R1628P increased risk, while R1398H had the opposite effect. G2019S increased the risk in European/West Asian and mixed populations while R1441C/G/H increased the risk of PD in European/West Asians. Combined with frequency analysis, we suggest that A419V, G2385R, and R1628P should receive top priority for screening in East Asian populations and that a greater focus be placed on G2019S and R1441C/G/H in European/West Asian populations.In conclusion, we found that LS, YZ, and BT: chose the topic and designed the experiments; LS, YZ, and QS: performed the analysis; LS, YZ, QS, and BT: analyzed the data; LS, YZ, and BT: wrote the manuscript; HP: data management and figure modification.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Dose-response analysis based on 5 case-control studies revealed that each 10 g/d increase in tofu intake was associated with 10% reduction in the risk of breast cancer . In summary, our findings suggest an inverse dose-response association between tofu intake and risk of breast cancer. However, owing to the limitations of case-control studies, more properly designed prospective studies are warranted to confirm this association.Observational studies on the association between tofu intake and breast cancer incidence have reported inconsistent results. We reviewed the current evidence and quantitatively assessed this association by conducting a dose-response meta-analysis. The electronic databases PubMed and EMBASE were searched for relevant studies published up to August, 2018. We included epidemiological studies that reported relative risks (RRs) or odds ratios (ORs) with 95% confidence intervals (CIs) for the association between tofu intake and breast cancer risk. A total of 14 studies  were included in the meta-analysis. The overall OR of breast cancer for highest vs lowest intake of tofu was 0.78 (95% CI 0.69–0.88), with moderate heterogeneity (P = 0.011, I Breast cancer is the most prevalent cancer and the leading cause of cancer death in females worldwide, with an estimated 1.7 million cases and 521,900 deaths in 2012 , 4. In 2Soy foods are traditionally more relevant to Asian diets than Western diets , 7. TheyThe meta-analysis was conducted in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) recommendation . The eleThe identification of relevant studies was performed independently by two different authors (QW and XL), and disagreements were resolved through consultation with a third reviewer (SR). Studies were included if they fulfilled the following criteria: 1) a case–control, cohort or cross-sectional design published in the English language; 2) studies evaluating the potential association between the risk of breast cancer and dietary tofu intake; 3) provided estimators of risk, such as relative risk (RR), incidence rate ratios (IRR), hazard ratios (HR), or odds ratios (OR) with 95% confidence intervals (CIs), or reported data to calculate them. The exclusion criteria were as follows: 1) experimental study; 2) letters or case reports; 3) articles that provided inadequate data or only information for breast cancer mortality. To avoid incorporating duplicated information, multiple publications from the same author or institution were seriously scrutinized, the study with the largest number of cases was included.The study characteristics were extracted independently by two authors (QW and SR) and any disagreement was resolved by consultation with a third reviewer (XL). From each article, the following variables were recorded: first author’s name, year of publication, study design, origin of the studied population, sample size, verification of breast cancer, range of categories of tofu intake, risk estimates with corresponding 95% CIs, adjustment for potentially confounding factors , and exposure assessment. Since the absolute risk of breast cancer is low, all the measures of association  are approximately equal to the estimates of OR, and the OR was used as the study outcome. When multiple ORs were available, we extracted estimates with the most comprehensive adjustment. When studies reported results separately by menopause or estrogen receptor (ER) status, both risk estimates were included separately in the analyses.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp). The range of possible scores is 0–9, and we considered studies which had a total NOS score of ≥7 as high quality for the meta-analysis.Quality assessment for each article was evaluated by according to the Newcastle Ottawa Scale (NOS), which was recommended by the Cochrane Non-Randomized Studies Methods Working Group . Therefore, we performed the meta-regression to explore the heterogeneity among studies . As a result, only the study quality (P = 0.003) was identified as a possible source of heterogeneity. Through the Galbraith plot, we noted that 2 studies [2 = 20.9%) after excluding the 2 studies, and the overall association was not materially changed .The estimated ORs for the highest versus lowest categories of tofu intake, from each study and all studies combined, are presented in  studies , 16, whiThe sensitivity analysis was also conducted by removing one study at a time and analyzing the remaining studies . The ORsBegg’s funnel plot and Egger’s test were used to evaluate the publication bias. The shape of the funnel plots showed obvious asymmetry , and theIn 2 = 40.8%).Next, we assessed the dose-response relationship between tofu intake and the risk of breast cancer, which included 5 case-control studies , 39, 40 There is relatively little epidemiological data available for the role of tofu in the etiology of breast cancer, in part because tofu is consumed mainly in East Asian populations. In the present analysis, evidence mainly from case-control studies are supportive of a protective role of tofu intake against breast cancer. The risk of breast cancer was reduced by 22% in a comparison of the highest with the lowest category of dietary tofu intake. This finding is consistent with the results from the previous meta-analysis by Qin et al.  in 2006,2 = 20.9%). In addition, the meta-regression analysis showed that the quality of the studies was a possible source of heterogeneity. When we stratified by study quality in the subgroup analyses, the association was weaker for studies of high quality  than low quality , with no evidence of study heterogeneity . These results suggested that several low-quality studies may have led to a lower risk estimate, although the combined ORs were still significant in the subgroup analyses for high-quality studies.Heterogeneity was detected in this meta-analysis in overall analysis. The Galbraith plot suggested that the observed heterogeneity seemed to be explained by Cho et al’s  and Kim Tofu is a versatile ingredient with many health benefits. As a particular soy food, It also contains high concentration of isoflavones, which possess both estrogen-dependent and -independent properties that potentially inhibit the development of breast cancer . It has Some limitations in our meta-analysis should be acknowledged. First, despite the strong inverse association between tofu intake and risk of breast cancer, our finding was based on two cohort studies and a large number of case-control studies, suggesting that our conclusion depend mainly on the case-control studies. Relatively speaking, case-control studies inevitably suffer some drawbacks like recall and selection biases, and thus our results should be interpreted with caution. Second, as a meta-analysis of observational studies, the potential impacts of unknown confounding factors on our findings cannot be completely excluded. For example, high tofu intake is characteristic of diets high in fruits and vegetables, low fat intake and increased physical activity, which are associated with lower breast cancer risk, although individual studies have considered a wide range of potential confounders in their analyses except for the study by Wu et al. , which hIn conclusion, evidence from case-control studies suggested an inverse dose-response association between tofu intake and breast cancer risk. Further well-designed prospective studies focusing on this association are necessary to confirm our findings.S1 PRISMA Checklist(DOC)Click here for additional data file.S1 Fig(TIF)Click here for additional data file.S2 Fig(TIF)Click here for additional data file.S3 Fig(TIF)Click here for additional data file.


Sixteen studies investigated the association between 25-hydroxyvitamin D and the risk of cancer mortality , and a higher 25-hydroxyvitamin D concentration was inversely associated with the risk of cancer mortality . Dose-response analysis indicated that the risk of cancer incidence was reduced by 7% , and the risk of cancer mortality was reduced by 2% , with each 20 nmol/L increment of 25-hydroxyvitamin D concentration. This meta-analysis provides evidence that a higher 25-hydroxyvitamin D concentration is associated with a lower cancer incidence and cancer mortality.Epidemiological studies have suggested inconclusive associations between 25-hydroxyvitamin D and total cancer incidence and mortality. The aim of this study was to quantitatively assess these associations by combining results from prospective cohort studies. A systematic literature search was implemented in PubMed and Scopus databases in April 2019. Comparing the highest with the lowest categories, the multivariate-adjusted relative risks (RRs) and the corresponding 95% confidence intervals (CIs) were pooled using a random-effects model. A trend estimation was performed using a two-stage, dose-response, meta-analysis method. Twenty-three independent prospective studies were included for data synthesis. Eight studies investigated the association between 25-hydroxyvitamin D and the risk of cancer incidence , and the summary estimate showed that 25-hydroxyvitamin D is marginally associated with cancer risk (Summary RR = 0.86; 95% CI: 0.73, 1.02; I Cancer is the leading cause of global morbidity and mortality, with approximately 18.1 million new cancer cases and 9.6 million cancer deaths in 2018 . In receAs a secosteroid hormone, the biologically active metabolite of vitamin D is attributed to 1,25-dihydroxyvitamin D 2D), which is synthesized from the circulating 25-hydroxyvitamin D (25(OH)D) by 1a-hydroxylase CYP27B1). The classical biological function of vitamin D is to regulate calcium and phosphorus balance and maintain bone health. Recent studies have found that it also plays an important role in anti-inflammatory, anti-fibrosis and immunoregulation processes, and it is also involved in maintaining the growth and development of the body and regulating cell proliferation, differentiation and apoptosis B1. The c,9. As eaIn fact, there were already several meta-analyses of prospective studies about cancer incidence or mortality, existing meta-analysis also yield inconsistent results ,27,28,29A systematic literature search was performed in the Cochrane library, Embase and PubMed databases until April, 2019 using the search query: Vitamin D, VD, 25-hydroxyvitamin D, 25(OH)D, 1,25-dihydroxyvitamin D or 1,25(OH)2D, paired with cancer, tumor, neoplasm, mortality, incidence, risk, and occurrence.Two authors independently performed the literature search and identified potential studies from the title, abstract and full-text. The inclusion criteria were prospective studies . The exposure of interest was the serum 25(OH)D concentration or plasma 25(OH)D concentration, and the end point of interestwas the number of cancer patients or deaths due to cancer during follow-ups.For the identified studies, data extraction was conducted by two investigators, and any discrepancy was resolved via discussion. The characteristics, including the first author’s surname, publication year, geographic region, duration of follow-up, mean age at the baseline, number of participants and cases, age of the participants, relative risk (RR), with the corresponding 95% confidence interval (CI), and the assay used to measure 25(OH)D, and covariates, adjusted for in the analyses, were extracted from the eligible studies. While most of the original studies used the lowest dose group as the reference group, a few used other groups as the reference group, so we converted this kind of study into the lowest dose group, as the reference group . The New2 statistic was used to estimate the heterogeneity between studies. To examine the potential sources of heterogeneity, subgroup and univariate meta-regression analyses, including duration of follow-up, region, study quality and mean age of participants, was implemented based on the information of these studies. In order to assess whether a study had an undue influence on the summary estimates, we deleted one study at a time and recalculated the summary estimates. The Egger’s test was used to assess publication bias D and cancer risk and mortality. Considering that the eligible studies were prospective studies, thus, odds ratio (OR) or HR was directly considered as RR for data synthesis. Multivariate-adjusted RRs, with the corresponding 95% CIs for the highest versus lowest category, were logarithm transformed, and the summary RR was calculated using a random-effects model, which was weighted by the inverse of the variance of the RRs. Studies including three or more 25(OH)D categories were eligible for dose-response analysis . The med   . Regardi= 0.012) . Dose-re= 0.012) , and the= 0.012) . 2 = 76.0%; p = 0.341), but there was no significant correlation in Europe [2 =0.002; p = 0.002). Studies were divided according to gender, and the male group [2 = 65.0%; p = 0.025), but there was no significant correlation in the female subgroup [2 = 75.8%; p = 0.042). The studies were stratified by the study quality, indicating that there was a borderline association between 25-hydroxyvitamin D and cancer risk in the high-quality group [In terms of total cancer incidence, subgroup analysis, stratified by region, indicated that the pooled effect of 25-hydroxyvitamin D was significantly associated with cancer incidence in Asia ,31 , but not in the male group [2 = 78.7%; p < 0.001). When the articles were divided into subgroups by the study quality, 25-hydroxyvitamin D was negatively correlated with cancer mortality in the high-quality subgroup [2 = 48.8%; p = 0.058), but there was no significant correlation in the moderate-quality subgroup [2 = 40.9%; p = 0.062). The studies stratified by region showed that there was a significantly association between vitamin D concentration and cancer mortality in Europe [2 = 38.1%; p = 0.104), and a marginal association in the US [2 = 48.8%; p = 0.012)  .In a sensitivity analysis, each study was sequentially deleted, and the remaining data were re-calculated. The results showed that the summary estimate was not substantially driven, to the exclusion of any one study. In publication bias analysis, non-significant publication bias was observed using Egger’s regression test .Analyses of 8 studies on cancer incidence and 16 studies on cancer mortality, showing that 25-hydroxyvitamin D is marginally associated with cancer risk and inversely associated with mortality, were pooled. The dose-response analysis showed that cancer risk decreased by 7% , with a 20 nmol/L increment of 25-hydroxyvitamin D, and cancer mortality decreased by 2% , with a 20 nmol/L increment of 25-hydroxyvitamin D. In fact, a meta-analysis of a prospective cohort study on the relationship between vitamin D concentration and cancer incidence and mortality was performed in 2013 . Based oThis study excluded the study of Giovannucci et al.  because As can be observed in our study, several studies have reported null associations between vitamin D concentration and cancer incidence or mortality ,47,51,54The inverse association between 25-hydroxyvitamin D and cancer incidence and mortality is biologically plausible. Existing studies have shown that vitamin D regulates the entire process of tumorigenesis, from initiation to stabilization, and the interaction with the cellular microenvironment ,23,62. VOur meta-analysis may have several limitations which must be taken into account. First, due to the observational nature of the data, it is possible that the observed significant inverse association between 25-hydroxyvitamin D and cancer incidence and mortality could be due to unmeasured or residual confounding. Second, the reduced cancer incidence and mortality may be associated with other health behaviors . A further limitation was that the eligible studies varied in several respects, including differences in the study populations, baseline comorbidities, and measurement of the exposure, and it is possible that methodological differences might have confounded the differences recorded across subgroups of the study. Not all prospective studies are conducted in the general population (e.g., the study of Pilz et al.  was condThe merits of this research should be emphasized. First, we have included more research than the previous meta-analyses. The large sample size of the meta-analysis provided powerful statistical power for the quantitative evaluation of this association. Second, eligible studies are medium- and high-quality prospective cohort studies that reduce the likelihood of recall errors and selection biases. Third, the confounding factors of the individual studies have been greatly adjusted to minimize biases. Besides, the follow-up of the prospective cohort studies is sufficiently long. Furthermore, dose-response analysis was also conducted to provide substantial evidence that vitamin D concentration is associated with cancer incidence and mortality, with a dose-dependent trend.The present study provides compelling evidence that a higher 25-hydroxyvitamin D concentration is marginally associated with cancer incidence and inversely associated with cancer mortality. Since the majority of the prospective studies were performed in Western countries, further prospective studies should be conducted in other regions, with different ethnic origins, to confirm these associations.


Objectives: The signal transducer and activator of transcription 4 (STAT4) gene encodes an important transcription factor that transmits signals induced by several cytokines associated with autoimmune diseases and has been identified as a susceptibility gene for numerous autoimmune disorders. The association between STAT4 rs7574865 polymorphism and the susceptibility of autoimmune thyroid disease (AITD) has been investigated in previous case-control studies. However, the investigation results were inconsistent. Hence, a meta-analysis was performed to draw a more reliable conclusion about it.Methods: All relevant studies were searched in Embase, PubMed, Web of Science, and China National Knowledge Infrastructure, till August 20, 2018. The pooled odds ratios (ORs) with 95% confidence intervals (CIs) were used to evaluate the strength of the association.Results: A total of five independent case-control studies with 1707 AITD patients and 2316 controls were included in the present meta-analysis. The overall pooled analysis indicated that STAT4 rs7574865 polymorphism was significantly associated with AITD susceptibility . However, the subgroup analysis showed a significant association of STAT4 rs7574865 polymorphism with AITD susceptibility in Asian population, but not in African population. STAT4 rs7574865 polymorphism was significantly associated both with Graves’ disease (GD) and Hashimoto’s thyroiditis (HT) susceptibility.Conclusion: This meta-analysis showed a significant association between STAT4 rs7574865 polymorphism and AITD susceptibility. However, further studies with larger sample sizes and other ethnicities are still required to confirm the findings. Autoimmune thyroid disease (AITD) affects approximately 5% of the general population . Graves’STAT4) gene, located on human chromosome 2q32.3, encodes a member of the STAT family of transcription factors. In response to cytokines and growth factors, STAT family members are phosphorylated by the receptor related kinases, and then form homo- or heterodimers that translocate to the cell nucleus in which they function as transcription activators  was the most common type of genetic variations in human genome. Some potentially functional SNPs have been reported to be associated with the risk of human complex diseases, such as cancer, schizophrenia, stroke, AITD . Thereinpression . Even thpression . For thele study . In ordeSTAT4 OR Signal transducer and activator of transcription 4,” “polymorphism OR SNP OR variation” and “autoimmune thyroid disease OR AITD”. In addition, the original or review reports were screened carefully and searched manually for more eligible studies based on their references.Embase, PubMed, Web of Science, and China National Knowledge Infrastructure (CNKI) electronic databases were searched for eligible studies published till August 20, 2018. The following terms were used for searching titles or abstracts: “STAT4 rs7574865 polymorphism and AITD susceptibility; (ii) studies that contained the distribution of allelic or genotypic frequencies of STAT4 rs7574865 polymorphism in all participants; (iii) studies that scored more than six stars based on the Newcastle-Ottawa Scale (NOS), which was adopted to assess the methodological quality of each candidate study (Eligible studies were chosen by two authors independently (XR Gao and JG Wang) according to the following inclusion criteria: (i) case-control studies that assessed the relationship between te study . The NOSP value for Hardy-Weinberg equilibrium (HWE). Any controversial content would be resolved by discussing with a third author (YG Yu).Two investigators (XR Gao and JG Wang) independently extracted data from each eligible literature, including first author’s name, publication year, country of origin, ethnicity, genotyping method, the total number of participants, allelic or genotypic frequencies, and PHWE value > 0.05 indicated that the genotype frequency of the controls was consistent with HWE. Chi-square based Q statistic were used to evaluate heterogeneity between studies. If PQ value < 0.1, the heterogeneity between studies was significant. Thus, the random-effects model would be adopted to calculate the pooled odds ratios (ORs), otherwise, the fixed-effects model would be chosen. The pooled ORs with 95% confidence intervals (CIs) were used to assess the strength of the association. The statistical significance of the strength was determined by Z-test. Pz value < 0.05 was considered statistically significant. A total of five genetic comparison models were used in the present analysis, including homozygote model (TT vs. GG), heterozygous model (TG vs. GG), dominant model [(TT+TG) vs. GG], recessive model [TT vs. (TG+GG)], and allele comparison model (T vs. G). In addition, the stability of the pooled results and the potential publication bias were evaluated by sensitivity analysis and funnel plots, respectively. All of the above analyses were performed by RevMan 5.0 software .Hardy-Weinberg equilibrium status of the controls was measured by the goodness-of-fit chi-square test. STAT4 rs7574865 polymorphism and AITD susceptibility. Thereinto, 1548 AITD patients and 2116 controls belonged to Asian population. Other paticipants including 159 AITD patients and 200 controls belonged to African population. For the NOS, all the studies scored more than 6 stars, and were considered to be of high methodological quality : OR = 1.55, 95%CI = 1.26–1.91, PZ < 0.0001]. The subgroup analysis based on ethnicity showed a significant association of STAT4 rs7574865 polymorphism with AITD susceptibility in Asian population , but not in African population. The stratified analysis by disease type showed that STAT4 rs7574865 polymorphism was significantly associated both with GD  and HT susceptibility .The association of Sensitivity analysis was conducted by removing a single study at a time and reanalyzing the pooled results. The results of the analysis indicated that the omission of any one study could not substantially change the overall pooled results under homozygote comparison (TT vs. GG), heterozygous comparison (TG vs. GG), and recessive comparison [TT vs. (TG+GG)] models. However, under dominant comparison [(TT+TG) vs. GG] and allele comparison (T vs. G) models, significant changes were observed after removing Hiz’s study , and the results indicated that the frequencies of STAT4 rs7574865 genotypes in GD patients were significantly different from that in the controls, and the T allele frequency of GD patients was also significantly higher than the controls. STAT4 rs7574865 polymorphism could act as a genetic risk factor of AITD. However, a case-control study  based on Tunisian population showed no significant association of STAT4 rs7574865 polymorphism with the risk of AITD (STAT4 rs7574865 polymorphism with AITD susceptibility, and observed that STAT4 rs7574865 polymorphism was significantly associated with AITD susceptibility. In addition, the subgroup analysis based on ethnicity showed a significant association of STAT4 rs7574865 polymorphism with AITD susceptibility in Asian population, but not in African population. The stratified analysis based on disease type indicated that STAT4 rs7574865 polymorphism was significantly related to the susceptibility of both GD and HT.In recent years, the association of tibility . Zhao et of AITD . Thus, iSTAT4 rs7574865 polymorphism and AITD susceptibility, but some existing limitations should not be ignored. Firstly, due to a limited number of related studies, we could not assess the association of rs7574865 polymorphism with AITD susceptibility among other ethnicities, such as Caucasians. Secondly, the sample size in the study was small, which reduced the clinical impact of the results. Therefore, further studies with larger sample sizes and other ethnicities are still required to confirm the current findings. Last but not least, our meta-analysis was based on an unadjusted assessment. A more precise analysis should be carried out if personal data such as gender, lifestyle and environmental exposure factors were available.The present results provided a stronger evidence for the association between STAT4 rs7574865 polymorphism and AITD susceptibility, especially in Asian population. The polymorphism may be used as a predictive marker for AITD predisposition.This meta-analysis showed a significant association between YY and XG designed the study. XG and JW searched the databases, collected the full-text papers, extracted and analyzed the data, and wrote the manuscript. YY reviewed the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


MTR) A2756G polymorphism on risk of developing pediatric acute lymphoblastic leukemia (ALL), but the available results were inconsistent. Therefore, a meta-analysis was conducted to derive a more precise estimation of the association between MTR A2756G polymorphism and genetic susceptibility to pediatric ALL. The PubMed, Embase, Google Scholar, Web of Science, ScienceDirect, Wanfang Databases and China National Knowledge Infrastructure were systematically searched to identify all the previous published studies exploring the relationship between MTR A2756G polymorphism and pediatric ALL risk. Odds ratios (ORs) and 95% confidence intervals (CIs) were applied to evaluate the strength of association. Sensitivity analysis and publication bias were also systematically assessed. This meta-analysis finally included ten available studies with 3224 ALL cases and 4077 matched controls. The results showed that there was significant association between MTR A2756G polymorphism and risk of pediatric ALL in overall population . In the stratification analyses by ethnicity, quality score and control source, significant association was found in Caucasians, population-based designed studies and studies assigned as high quality. In conclusion, this meta-analysis suggests that MTR A2756G polymorphism may influence the development risk of pediatric ALL in Caucasians. Future large scale and well-designed studies are required to validate our findings.Plenty of studies have investigated the effect of methionine synthase ( Acute lymphoblastic leukemia (ALL) is the most common childhood cancer, which accounts for 30% of all malignancy diagnosed in children and 80% of pediatric leukemia . However12-dependent enzyme, which remethylates homocysteine to methionine and simultaneously generates tetrahydrofolate by removing methyl group from 5-methyltetrahydrofolate. MTR helps to maintain the levels of adequate intracellular folate and normal homocysteine and methionine concentrations, which are used for proper DNA methylation or other methylation processes [MTR gene is mapped on 1q43, and the extensively investigated A2756G polymorphism (rs1805087) leads to a change from aspartate to glycine at codon 919 (D919G), resulting in reduced enzyme activity [MTR A2756G polymorphism has been linked to various cancer, such as prostate cancer, retinoblastoma and lymphoma [MTR A2756G polymorphism on pediatric ALL risk, yet the reported results are inconsistent. The inconsistencies of results might be attributed to some variables in study of population like genetic backgrounds difference and relatively small size of sampling in single study. Therefore, a meta-analysis was conducted to derive a more precise estimation of the association between MTR A2756G polymorphism and genetic susceptibility to pediatric ALL.The methionine synthase (MTR) plays a crucial role in the folate metabolic network. It is a vitamin Brocesses . The MTRactivity . It has activity ,16. Plenlymphoma . A numbeMTR A2756G polymorphism and pediatric ALL risk with the following subject terms or keywords: ‘methionine synthase’ or ‘MTR’ or ‘MS’ or ‘5-methyltetrahydrofolate-homocysteine methyltransferase’, ‘polymorphism’ or ‘variation’ or ‘variant’ or ‘mutation’, ‘acute lymphoblastic leukemia’ or ‘leukemia’ or ‘ALL’, and ‘pediatric’ or ‘children’ or ‘childhood’. The latest web-based literature search was conducted on May 20, 2018 and no language restriction was applied. In addition, the reference lists in the primary studies and review articles were also examined manually to identify additional potentially relevant studies.The present study was conducted in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines. The PubMed, Embase, Google Scholar, Web of Science, ScienceDirect, Wanfang Databases and China National Knowledge Infrastructure were systematically searched to identify the published case–control studies on the relationship between The following inclusion criteria were applied for selecting literature: (1) confirmed diagnosis for the pediatric ALL cases; (2) case–control study; (3) available genotypes distribution data for both patients and control populations; (4) genotypes distribution of the control group must be in consistent with Hardy–Weinberg equilibrium (HWE). The case reports, letters, commentary and review articles were excluded. If the same or overlapping patient population was reported by several articles, only the most recent or largest sample size was chose in this meta-analysis.The quality assessment of included studies was preformed independently by two authors according to the Newcastle-Ottawa Scale (NOS). Discrepancies were adjudicated by the third investigator until consensus was achieved. The NOS is a tool used for assessing the quality of non-randomized studies included in a systematic review and meta-analysis . Using tMTR A2756G polymorphism in cases and controls. Any disagreement was resolved by re-evaluation of the originally included studies.From each of the included articles, the following data were collected independently by two authors: the name of first author, year of publication, country and ethnicity of participants, source of controls, total number of cases and controls, genotyping methods, genotyping data of the 2χ-test was adopted to check HWE of genotypes distribution frequencies, with P<0.05 indicating deviation from HWE. The strength of association between MTR A2756G polymorphism and pediatric ALL risk was assessed by calculating pooled ORs and corresponding 95% CIs under the allele model , heterozygote model (AG vs. AA), homozygote model (GG vs. AA), recessive model (GG vs. AA+AG) and dominant model (AG+GG vs. AA), respectively. The significance of the overall ORs was determined by the Z-test. The 2χ-test based Q-test was performed to estimate the heterogeneity across the eligible studies, and the heterogeneity was further quantified with I2-test. When P>0.05, showing that the effects were assumed to be homogeneous, the fixed-effects model (Mantel–Haenszel method) was selected to calculate the ORs, alternatively, the random-effects model (DerSimonian–Laird method) was used [P values were two-sided, and P<0.05 was considered statistically significant.For the controls of each study, the was used . Stratifwas used . All theMTR A2756G polymorphism in cases and controls from the individual studies were shown in The flow diagram of literature selection was presented in MTR A2756G polymorphism and risk of pediatric ALL under three genetic models  ( = 0.03) . In the  = 0.03) , Table 3MTR A2756G polymorphism, so the fixed-effects model based Mantel–Haenszel method was selected for the combined analysis , and homozygosity (GG) with a 1.88-fold increased risk of ALL . de Jonge et al. [MTR A2756G polymorphism between children ALL and the controls. To elucidate this inconsistency, a meta-analysis was conducted to derive a more precise estimation of the association.Numerous investigations have examined the association of a et al.  found the et al.  found noMTR A2756G polymorphism and risk of pediatric ALL in overall comparison. Individuals with the MTR 2756 G allele had increased risk of developing pediatric ALL compared to those with the A allele. Moreover, individuals with the AG genotype or the AG+GG genotype had raised risk of pediatric ALL compared to those with the AA genotype. Significant association was also found in Caucasians, population-based designed studies, and studies assigned as high quality. Our results were in accordance with the conclusion reported by Xia et al. [MTR 2756 A allele was associated with a decreased risk of pediatric ALL compared with the G allele. In present meta-analysis, more web-based databases including English and non-English databases were systematically searched to minimize the selection bias and the potential risk of missing eligible literature [In the present study, the combined results found that there was significant association between a et al. , which sterature . Since oMTR 2756AG individuals who were SHMT1 1420CT/TT had a 5.6-fold reduction in ALL risk [MTR 2756 G was a risk allele for ALL on itself but also in combination with the MTHFR 677 T allele in adults [MTR A2756G polymorphism is somewhat diluted or concealed by other gene–gene interactions. Future studies combining other genes in folate metabolism with MTR are encouraged.When interpreting the results, some limitations of our meta-analysis should be considered. First, our results were based on unadjusted estimates, which may cause confounding bias. A more precise analysis could be performed if all raw data were available, which would allow for the adjustment by other confounders including sex, age, lifestyles and other potential factors. Second, the quantitative synthesis of some subgroups may have no sufficient testing power to accurately assess the real association, for instance, only one study was conducted among Asians. In addition, the gene–environment interactions which may modify genetic susceptibility to cancer were not taken into account in the present study due to the limited data. Last but not least, we also did not consider other genes in folate metabolic network that might be associated with the risk of pediatric ALL. The etiological mechanism of ALL is very complicated, in which gene–gene, and gene–environment interactions are involved ,38. SeveALL risk . In contMTR A2756G polymorphism influences the genetic susceptibility to pediatric ALL, especially in Caucasians. However, large scale and well-designed studies are required to validate our findings, and the biochemical mechanism and function of MTR A2756G polymorphism should also be investigated in the future.In conclusion, this meta-analysis suggests that


P=0.039; AA vs. AG+GG: OR & 95% CI, 1.93, P=0.049) or A allele carriers , P=0.047) of the rs3850641 polymorphism of the TNFSF4 gene increased the risk of MI. In conclusion, this case–control study confirms that the rs3850641 polymorphism of the TNFSF4 gene increases the risk of MI.Tumor necrosis factor superfamily member 4 (TNFSF4), also known as Ox40 ligand (Ox40l), plays an important role in atherosclerosis development. Several studies reported the association between the rs3850641 polymorphism of the TNFSF4 gene and the risk of myocardial infarction (MI). However, the results are inconsistent. In order to explore the relationship between the rs3850641 polymorphism of the TNFSF4 gene and MI, we conducted a case–control study including 454 cases and 512 controls in a Chinese Han population. Genotyping was performed using matrix-assisted laser desorption/ionization time-of-flight mass spectrometry. The present study found that AA genotype (AA vs. GG: odds ratio (OR) & 95% confidence interval (CI), 2.00,  Coronary heart disease (CHD) is a significant risk factor for mortality, with myocardial infarction (MI) being the most serious  consequence of CHD . The maiTumor necrosis factor superfamily number 4 (TNFSF4), also known as OX40 ligand (OX40L), is located in human chromosome 1 and encodes a type II glycoprotein. The expression of TNFSF4 has been observed in T cells, B lymphocytes, vascular endothelial cells, macrophages, mast cells, and smooth muscle cells, all of which are involved in the development of atherosclerosis ,7. MalarSeveral studies have reported an association between the rs3850641 polymorphism of the TNFSF4 gene and the risk of MI . HoweverA hospital-based case–control design was used in the present study. In total, 454 hospitalized MI patients were recruited from the First Affiliated Hospital Zhengzhou University between January 2013 and May 2017. Diagnosis of MI was confirmed by identifying stenosis in any of the major coronary arteries or in the left main trunk using coronary angiography.In total, 512 controls were randomly selected among subjects who received regular health examinations at the First Affiliated Hospital Zhengzhou University between January 2013 and May 2017. Individuals with congestive heart failure, peripheral vascular disease, rheumatic heart disease, pulmonary heart disease, chronic kidney disease, hepatic disease, or any malignancy were excluded from the study.Demographic, lifestyle, and clinical characteristics of all patients and controls, including age, body mass index (BMI), total cholesterol (TC), low-density lipoprotein (LDL), and high-density lipoprotein (HDL) were extracted from medical records. Written informed consent was obtained from all included patients and controls. We obtained approval for the study protocol from the Ethics Committee of the First Affiliated Hospital Zhengzhou University. The ethical approval of our study was in line with the standards of the Declaration of Helsinki.Blood samples were collected using vacutainer tubes and then transferred to ethylenediaminetetraacetic acid (EDTA) tubes. Genomic DNA was isolated from whole blood using the QIAamp DNA Blood Mini Kit . Genotyping was performed by matrix-assisted laser desorption/ionization time-of-flight mass spectrometry (MALDI-TOF MS) as previously described . SNP gen2) test  and Student’s t-test (for continuous variables). The associations between the rs3850641 polymorphism A/G genotypes of the TNFSF4 gene and the risk of RA were estimated by calculating odds ratios (ORs) and 95% confidence intervals (CIs) using logistic regression analysis, and by crude ORs and adjusted ORs when adjusting for age and sex. The Hardy–Weinberg equilibrium (HWE) was assessed by a goodness-of-fit χ2 test to compare the observed genotype frequencies with the expected frequencies in controls. The present study was powered to detect the effect of the rs3850641 polymorphism of the TNFSF4 gene on MI susceptibility at a P value of 0.05 [Demographic characteristics and rs3850641 polymorphism genotypes of the TNFSF4 gene were evaluated using a chi-squared   . HoweverP=0.047) . We furtP=0.047) .To investigate whether the power of the present study was sufficient given the sample size, we calculated the power of the rs3850641 polymorphism in the allelic model. Analysis showed that our study had a power of 72.1% to detect the effects of the rs3850641 polymorphism on MI susceptibility, assuming an OR of 1.27.In the present study, we investigated the association between the rs3850641 polymorphism of the TNFSF4 gene and the risk of MI in a Chinese population and found that the polymorphism conferred an increased risk of MI.TNFSF4 is a T-cell activating factor that seems to facilitate the survival and/or promote anti-CD3-induced CD4+ T cells proliferation at the time of inflammation . T cellsIn the stratification analyses, we found that the risk of MI conferred by the rs3850641 polymorphism of the TNFSF4 gene remained significant in the male, nonsmoking, and nondrinking subgroups. This can be explained by the concept that susceptible individuals are likely to have a degree of exposure to risk factors. However, given the decreased sample sizes in the stratification analyses and the limited power, the results should be interpreted with caution. Despite this, our findings still provide evidence for a possible interaction between the SNP and certain MI risk factors.Several potential limitations of this case–control study should be considered. First, the patients and controls were recruited from hospitals and may not be representative of the general population. Nonetheless, the genotype distribution of the controls was in HWE. Second, a single case–control study may not be sufficient to fully interpret the relationship between the rs3850641 polymorphism of the TNFSF4 gene and susceptibility to MI, because of the limited sample size. Larger numbers of subjects are necessary to confirm our findings. Third, we did not obtain detailed information about MI severity and response to treatment, which restricted our analyses. Fourth, the risk of MI cannot be attributed to a single TNFSF4 gene SNP; other SNPs in the TNFSF4 and other genes, as well as certain environmental factors, should also be considered. Fifth, the underlying mechanisms of this SNP in MI should also be investigated. Finally, further studies in different population may establish the true significance of the association between this SNP and MI risk.In conclusion, our study provides strong evidence that the rs3850641 polymorphism of the TNFSF4 gene may contribute to MI risk. However, our results were obtained in a sample of limited size, and this therefore represents a preliminary conclusion. Validation through multicenter case–control studies with diverse ethnic populations is needed to confirm our finding.


With the exception of renal cell carcinoma, studies assessing the association between hypertension and other cancers are inconsistent. We conducted a meta-analysis to assess this evidence. We included observational studies investigating the association between any definition of hypertension or systolic and diastolic blood pressure and risk of any cancer, after searching PubMed until November 2017. We calculated summary relative risks (RR) and 95% confidence intervals (CI) using inverse-variance weighted random effects methods. A total of 148 eligible publications were identified out of 39,891 initially screened citations. Considering only evidence from 85 prospective studies, positive associations were observed between hypertension and kidney, colorectal and breast cancer. Positive associations between hypertension and risk of oesophageal adenocarcinoma and squamous cell carcinoma, liver and endometrial cancer were also observed, but the majority of studies did not perform comprehensive multivariable adjustments. Systolic and diastolic blood pressure were positively associated with risk of kidney cancer but not with other cancers. In addition to the previously well-described association between hypertension and risk of kidney cancer, the current meta-analysis suggested that hypertensive individuals may also be at higher risk of colorectal and breast cancer. However, careful interpretation is required as most meta-analyses included relatively small number of studies, several relative risks had weak or moderate magnitude and maybe affected by residual confounding. It is responsible for almost 50% of heart disease, stroke, and heart failure and about 14% of total deaths in 2015 were related to hypertension2. In addition, an estimated 10% of the healthcare spending is directly related to increased blood pressure and its complications3.Hypertension and cancer are two multifactorial, severe and chronic conditions. Hypertension is a major health problem worldwide, as it affects approximately 3 in 10 adults over age 20, leading to high morbidity and mortality4. Therefore, if blood pressure and/or hypertension are associated with the risk of cancer, this may have important public health consequences.Cancer is also a leading cause of morbidity and mortality. Worldwide, there were approximately 18 million new cases and 9.6 million cancer-related deaths in 2018, and these numbers are expected to rise within the next two decades6, although it is not clear yet whether reverse causation explains part or all of this association7. Claims of association also exist for colorectal, breast, endometrial and prostate cancer12, but the evidence is inconsistent. But recently there has been an influx of new studies investigating the association of metabolic syndrome, which includes hypertension, and cancer risk14, and therefore we conducted a systematic review and meta-analysis to summarize and evaluate this literature.Renal cell carcinoma is the only cancer type that has been consistently associated with hypertension(blood pressure OR hypertension OR “metabolic syndrome”) AND , and also hand-searched for potential missed studies in the references of eligible systematic or narrative reviews24. We considered all cohort and case-control studies conducted in humans assessing the association between systolic blood pressure (SBP) or diastolic blood pressure (DBP) or hypertension and risk of any cancer development or cancer death. Hypertension was defined as any of the following: self-reported or measured SBP or DBP higher than predefined cut-off points (e.g. WHO or NCEP/ATP III criteria), self-reported medical history of hypertension or drug-treated hypertension. When the exposure was metabolic syndrome, we only used the hypertension component. We excluded narrative or systematic reviews, case-series or case-reports, cross-sectional studies, prognostic studies among cancer patients, studies in children, cohort studies with a sample size of less than 100 total individuals, case-control studies with less than 100 cancer cases, studies that did not provide enough information for calculating measures of association, studies in other than the English language and studies of pre-malignant outcomes . Whenever published reports pertained to the same cohort evaluated at different follow-up periods, we retained the publication with the longer follow-up.We searched for eligible published studies in PubMed from inception to November 2017 using the following search algorithm: We recorded information from each eligible study on author name, year and journal of publication, study design, sample size, population characteristics, definition of hypertension, relative risk (RR), 95% confidence interval (CI) and adjustments for confounders. When results from multiple statistical models were reported, we always retained the most adjusted model. The literature search and data extraction were performed independently by two investigators (AS and XT), and data was re-checked for consistency by a third author (SC).25.We used the “Newcastle-Ottawa Scale” to assess the quality of the included studies, and this task was performed independently by two investigators (AS and SC), and disagreements were resolved by a third author (KKT). Based on this tool, every study was judged on eight quality items, grouped into three categories: i) selection, ii) comparability of study groups and iii) exposure or outcome ascertainment for case-control and cohort studies, respectively. A star was awarded for every quality item for a maximum of 9 stars for the highest quality studies27. We summarized RRs and 95% CIs using fixed-effects and random-effects meta-analysis models, if three or more studies were available per exposure and outcome comparison28. Between-study heterogeneity was assessed by the Cochran’s Q test and the I2 statistic29. 95% prediction intervals were also calculated to further assess heterogeneity, which represents the range in which the effect estimates of future studies will lie30. Subgroup meta-analyses were conducted according to sex, study design (prospective vs. case-control) and adjustment factors (at least age vs. age plus further multivariable adjustment). Presence of small-study effects was assessed using the Egger’s regression asymmetry test31. Analyses were performed in STATA 12 . All p-values were two-tailed. The study is reported according to the MOOSE checklist32.Our primary analysis included only prospective studies , whereas analyses including case-control studies are provided in the online supplement. We conducted separate meta-analyses according to systolic or diastolic blood pressure and hypertension. Analyses for systolic and diastolic blood pressure used both a top to bottom category comparison approach and a dose-response approach per 10 mmHg. Studies that reported a dose-response estimate were either pooled directly or after re-scaling to the corresponding increment unit using the generalized weighted least-squares regression model approach116, 72 of which were cohort studies, 11 were nested case-control studies111 and 2 were record-linkage studies79, and 63 case-control studies178. Most studies  investigated associations with cancer incidence. A total of 48 (32%) studies were conducted only in men, whereas 57 (39%) studies were conducted only in women. The majority of the studies  were conducted in the USA followed by Scandinavian countries , Italy , UK and Korea , and finally Japan with 6 studies (4%). Out of the 128 studies that used hypertension as the exposure of interest, this was defined in 20 studies (16%) using the NCEP-ATPIII criteria (≥130/85 mmHg), 17 studies (13%) used the WHO definition (≥140/90 mmHg), 4 studies (3%) used ≥160/95 mmHg, 38 studies (30%) pertained to self-reported hypertension, 16 studies (13%) used self-reported drug treatment for hypertension and 26 studies (20%) used a combination of self-reported disease and treatment, whereas 7 studies used other definitions.Figure Detailed information on the quality assessment for each study, using the Newcastle-Ottawa scale, can be found on Supplementary Table For the primary analysis using only prospective studies, we conducted 30 meta-analyses of hypertension and 16 different cancers, and 29 meta-analyses between SBP or DBP and 11 cancers. Summary random effects relative risk estimates and 95% CIs per cancer site, heterogeneity statistics, 95% prediction intervals and tests of small-study effects are summarized in Fig. 109. We observed large heterogeneity , but no indication for small study effects 179. The association between hypertension and kidney cancer was statistically significant in both women  and men  without between-study heterogeneity within each sex-specific analysis, but considerable larger estimates were observed in women . Similar results were obtained in subgroup meta-analyses of prospective studies that at least adjusted for age vs. studies that adjusted for age plus at least three of the following five risk factors: body mass index, smoking, alcohol, physical activity and family history of kidney cancer  was observed  and DBP  with kidney cancer, but these meta-analyses had evidence of small-study effects , an association that remained significant after excluding one record-linkage study 76 or after meta-analysing only the four studies that performed multivariable adjustments 114. This association was statistically significant in men 114, but not in women 113. No significant associations were identified between SBP or DBP with colorectal cancer risk 110, but the association was not significant for post-menopausal breast cancer risk 87. However, statistically significant associations were observed for total 85, and post-menopausal  breast cancer risk in prospective studies that performed multivariable adjustment  and post-menopausal disease .The meta-analysis yielded a borderline significant association between hypertension and risk of breast cancer . After removing two record linkage studies79, we observed a significant RR of 1.37  but again the heterogeneity was substantial . Only two out of the nine prospective studies used multivariable adjusted models, and the meta-analysis among them yielded null results 116. When 13 case-control studies were meta-analysed together with the prospective studies, a statistically significant association was found . No analysis on SBP or DBP was performed, because there were not enough studies using continuous BP data.Our meta-analysis showed also a statistically significant association between hypertension and risk of endometrial cancer , but four out the five studies did not perform comprehensive multi-variable adjustments 111. We identified only 2 prospective studies on hypertension and esophageal squamous cell carcinoma77, and the meta-analysis yielded a statistically significant estimate , but was based only on 248 cases. Finally, the meta-analysis of all 6 prospective studies on total oesophageal cancer yielded a statistically not significant estimate . Most of these studies however did not perform comprehensive multivariable adjustments . Furthermore, a dose-response approach revealed a 5% and 7% higher risk for kidney cancer per every 10 mmHg higher SBP and DBP, respectively. Similar associations have been reported in previous meta-analyseset al. published another meta-analysis in 2013, which found a 9% relative increase in colorectal cancer due to hypertension using both cohort and case-control studies10. To our knowledge, no clear mechanism has been proposed to link hypertension to colorectal cancer, but hypertension has been shown to increase cancer risk by blocking apoptosis187. Current findings should be approached with caution given the scarcity of well-designed studies in the literature.We also estimated a positive summary association between hypertension and risk of colorectal cancer with a higher risk of 11% for hypertensive individuals. This meta-analysis included data from 13 prospective studies, only four of which performed comprehensive multivariable adjustments, and the potential higher risk was 30% for hypertensive individuals in these studies. Esposito et al. that reported a 7% higher risk for total breast risk in hypertensive individuals regardless of performed adjustments11. Suggestive mechanisms to explain this association involve blocking of apoptosis, adipose tissue related hypoxia and chronic inflammation promoting reactive oxygen species formation188.In a meta-analysis of 13 prospective studies, hypertension was associated with a 7% higher risk of total breast cancer. Only five of these studies performed comprehensive multivariable adjustments, where the risk was 10% for total breast cancer and 38% for post-menopausal disease comparing hypertensive to normotensive individuals. These findings are in agreement with a recent meta-analysis of 12 prospective studies by Han 9.Positive associations between hypertension and risk of esophageal adenocarcinoma and squamous cell carcinoma, liver, and endometrial cancer were also observed in the current meta-analysis, but these meta-analyses included a small number of prospective studies, ranging from 2 for oesophageal squamous cell carcinoma to 9 for endometrial cancer, and the majority of them did not perform comprehensive multivariable adjustments, raising serious concern over the validity of the estimated associations. Previous meta-analyses exist for endometrial cancer, but have in general shaped their conclusions without taking into consideration study design and quality12.We did not observe associations between hypertension or blood pressure and risk of stomach, gallbladder, pancreas, lung, cervix, prostate, bladder and brain cancer. A recent meta-analysis examining the association between hypertension and risk of prostate cancer used 21 cohort and case-control studies and found a statistically significant 8% higher risk, but again this report shaped its conclusions without taking into consideration individual study design and qualityStrengths of the present meta-analysis include the comprehensive search strategy with the inclusion of “metabolic syndrome” in the search algorithm, the wide scope of investigating associations with many individual cancers and the detailed sensitivity and subgroup analyses. There are also limitations in this work. First, the association between hypertension and malignancy may be due to shared risk factors such as age, smoking, alcohol consumption, diet and adiposity. Unfortunately, the majority of the literature in this field contains studies that did not perform comprehensive multivariable adjustments, which raises concerns over the robustness of individual study findings even for the well-described association between hypertension and kidney cancer. Second, most studies did not perform subgroup analyses by potential effect modifiers (e.g. anti-hypertensive medication use), which would have allowed a more accurate assessment of associations in different patient subgroups. Third, detection bias may also account for some of the reported associations, as individuals treated for hypertension are under closer medical surveillance that may lead to easier detection of cancer compared to untreated persons. Future well-designed prospective studies and Mendelian randomization studies should assist in estimating valid measures of association.In conclusion, the currently available published observational evidence across 18 cancer sites showed that hypertensive individuals were at higher risk of kidney, colorectal and breast cancer. However, careful interpretation is required as most meta-analyses included relatively small number of studies, several relative risks had weak or moderate magnitude and maybe affected by residual confounding.Supplementary Information


Objective: Both tooth loss and dementia are age-related and frequently-occurring diseases. Increasing attention has been given to explore the pathogenesis related to oral-brain function disorders. The present study was performed to evaluate the association between tooth loss and dementia through a dose-response meta-analysis.Methods: Relevant cohort studies were searched from online databases up until June 20, 2018, which examined the association between tooth loss and the risk of dementia. Literature selection according to inclusion and exclusion criteria, as well as data extraction from included studies were completed independently by two reviewers. Data syntheses in this meta-analysis were performed using Stata 12.0 software.Results: A total of 8 cohort studies were included, containing a total of 14,362 samples and 2,072 dementia patients. The result of the meta-analysis indicated that patients with tooth loss faced a 1.34 times greater risk of developing dementia . The result from this dose-response meta-analysis in a linear model, suggested that every missed tooth might increase the risk of dementia by 1.01 times . Further subgroup analyses pointed out that tooth loss patients without dentures may have a higher risk of dementia than those with dentures ; at the same time, the study design, study area and education level of the study participants, might also have some effect on the results.Conclusions: Tooth loss may be a risk factor for the development of dementia. In addition, there is a dose-response relationship with the increase of missing teeth. Tooth loss frequently occurs in patients with dental caries, periodontal diseases, tumors, traumatism and retrogression, accompanied by masticatory dysfunction, hypotrophic absorption, pronunciation difficulty and anile appearance. Nowadays, tooth loss is no longer regarded as a simple local disorder, but also as a risk factor for systemic diseases, such as cardiovascular diseases  as the most common type and accounts for 60–70% of all dementia cases Mayeux, . The maiOur study conformed to the recommended Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement ,” eligible studies were selected based on the following inclusion criteria: (1) interested population consisted of patients suffering tooth loss and/or dementia without restrictions on gender, age and race; (2) interested exposure was tooth loss; (3) participants in the control group had no tooth loss; (4) interested outcome was referred to dementia; (5) study design restrictively followed cohort design; and (6) with full text and relevant data as follows: hazard ratio (HR), odds ratio (OR), relative risk (RR) with 95% confidential interval (CI) or raw data for calculating crude HR, OR, or RR with corresponding 95% CI. If outcomes were duplicated or shared in more than one study from one institution, the most comprehensive study was chosen for the analysis.Studies would be excluded if they conformed to any one of the following conditions: (1) animal experiments; (2) focusing on risk factors (such as periodontitis and caries) for tooth loss but without risk estimation on tooth loss; (3) interested outcomes could only be classified into clinical manifestations of dementia , but did not conform to dementia itself.We searched the databases of PubMed, Embase, China National Knowledge Infrastructure (CNKI) and the International Clinical Trial Registry Platform (ICTRP) for relevant cohort studies published before June 20, 2018, and the references of relevant papers were also examined to detect additional studies. The following items were applied in literature searching: “tooth loss” or “edentulous” or “edentulism” or “periodontal disease” combined with “dementia” or “Alzheimer's disease” or “risk factor.”The studies finally included were selected according to inclusion and exclusion criteria. A designed form was then utilized to collect the following information from the included studies: the name of the first author, year of publication, the type of study design, study area, sample size, participants' age, follow-up duration, cognitive assessment, tooth loss assessment, risk estimation results , and adjustment for covariates. Both the study selection and data extraction were conducted by two reviewers independently, cross checked and any disagreements over the data between the reviewers, were resolved through discussion.I2 statistics and a Q test, and the fixed-effect model was utilized when I2 ≤ 50% and p > 0.1; otherwise the random-effect model was preferred , and determined the adoption of the fixed-effect model for analysis. The analysis result indicated that patients with tooth loss had a 1.34 times higher risk of developing dementia , as shown in Figure The heterogeneity detected between the 11 datasets, was low (P = 0.52) and the dose-response meta-analysis suggested that every missing tooth might increase the risk of dementia by 1.01 times  in a linear model, with the risk increasing to 1.18 times, when the number of missing teeth went up to 20 . Correspondingly, edentulous cases faced a 1.31 times higher risk of dementia .A further analysis investigated the dose-response relationship of the two diseases based on six studies with eight datasets .The symmetry in Figure Our present study investigated the association between tooth loss and dementia based on eight cohort studies involving a total of 14,362 cases and 2,072 dementia patients. The result of this meta-analysis indicated that tooth loss statistically increased the risk of dementia by 1.34-fold. A further dose-response analysis detected a linearity association between the two diseases, suggesting that the more teeth are missing, the higher the risk of dementia. In subgroup analyses, we found that the association between the two illnesses might be affected by certain factors, such as study design, study area, denture status and education level, but was not affected by follow-up duration, risk expression, the strength of confounding factor adjustments, or adjustments for smoking, alcohol and cardiovascular and cerebrovascular diseases.A positive association between the two diseases, detected in tooth loss patients without dentures, might be partially explained by the alteration of chewing ability  and the lack of significant publication bias (P = 0.29).Apart from the present meta-analysis, three previously published meta-analyses have explored the association between tooth loss/ periodontitis and dementia, and revealed that tooth loss might be a risk factor for the development of dementia (Shen et al., However, there were some limitations in this study. First, only eight studies meeting inclusion criteria were included into the meta-analysis. Second, it may not be appropriate to pool HRs and ORs directly without data conversion, though the subgroup analysis indicated that the overall result was not significantly affected by the different risk expressions. Third, our study failed to explore the association of different types of dementia, because most studies did not specifically report the subtypes of dementia, with only one study separately stating the subtypes of Alzheimer's disease and vascular dementia (Takeuchi et al., In conclusion, the results from this study indicated that tooth loss may be a risk factor for dementia, and dementia risk may increase with the increase in the number of teeth los. In addition, timely prosthodontic treatment with dentures may reduce the risk of dementia related to tooth loss. Therefore, in clinical practice, tooth loss should be regarded as a marker for some systemic disorders, including dementia. Moreover, necessary intervention should be adopted for tooth loss. However, it remains to be determined whether removable dentures can be safely employed in tooth loss patients with dementia. Internal connections of chewing ability, nutrient intakes, increased inflammatory markers, and psychosocial factors, as well as clarifying the susceptibility among tooth loss patients to the various subtypes of dementia, should be further explored.JC, W-DL, and X-TZ designed the study. C-JR and L-YX performed the data searches and collected the data, JS re-checked the data. LW and X-TZ performed the analysis. JC and LW wrote the manuscript, W-DL and X-TZ reviewed the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Background: Non-syndromic cleft lip/palate (NSCL/P) has an etiology, including both genetic and environmental factors. Herein, we evaluated the association of rs13041247 and rs11696257 v-maf musculoaponeurotic fibrosarcoma oncogene homolog B (MAFB) polymorphisms with the risk of NSCL/P in a meta-analysis. Methods: The PubMed/Medline, Scopus, Cochrane Library, Web of Science, and HuGE Navigator databases were systematically searched to retrieve relevant articles published up to January 2019. The Newcastle–Ottawa scale was applied for quality evaluation of retrieved articles. The 95% confidence interval (CI) and crude odds ratio (OR) were calculated for each study using the Review Manager 5.3 software to show the association between MAFB polymorphisms and risk of NSCL/P. The comprehensive meta-analysis 2.0 software was used to calculate the publication bias. In addition, sensitivity analysis was carried out to show the stability of results. Results: Of 102 articles retrieved from the databases, 10 articles were analyzed in this meta-analysis. Ten articles, including eleven studies reporting rs13041247 MAFB polymorphism, included 3082 NSCL/P patients and 4104 controls. Three studies that reported rs11696257 MAFB polymorphism involved 845 NSCL/P patients and 927 controls. The rs11696257 MAFB polymorphism was not associated with the risk of NSCL/P, but the CC and TC genotypes of rs13041247 polymorphism were associated with the risk of NSCL/P. Nevertheless, the C allele and CC and TC genotypes were associated with a significant decline in the risk of NSCL/P in population-based studies. Conclusions: The results of this meta-analysis demonstrated that the risk of NSCL/P was related to rs13041247 polymorphism, not rs11696257 MAFB polymorphism. Well-designed studies are required to assess the interaction of MAFB and other genes with environmental factors in different ethnic groups. MAFB) encoded by the MAFB gene [MAF family of transcription factors characterized by a typical basic leucine zipper (bZip) structure [MAFB polymorphisms in the development of NSCL/P in the Asian population [MAFB polymorphisms with the risk of NSCL/P.The interaction between genetic and environmental factors may result in the development of non-syndromic cleft lip/palate (NSCL/P), which is the most common human congenital anomaly worldwide ,2. This AFB gene . This geAFB gene  as well AFB gene ,11,12. Ttructure ,11,12 thtructure . Assessmtructure . In addipulation . TherefoThis meta-analysis is reported according to the PRISMA guidance .MAFB or v-maf musculoaponeurotic fibrosarcoma oncogene homolog B) and  with no language restriction. In addition, we investigated the references of the retrieved articles related to the topic to make sure no article was missed.The PubMed/Medline, Scopus, Cochrane Library, Web of Science, and HuGE Navigator databases were systematically searched to retrieve relevant articles published up to January 2019. The searched terms were: (The first author (M.S) searched for articles in the databases and evaluated the titles and abstracts of the relevant articles, and then, the full-texts of the articles that met the eligibility criteria were uploaded and screened. The reason for the exclusion of any article excluded after full-text screening was reported. The second author (M.M.I) independently re-assessed the relevant articles. If there was a disagreement between the two authors, the problem was resolved by discussion.MAFB polymorphisms, and (4) studies having the sufficient data to calculate the odds ratios (ORs) and 95% confidence intervals (CIs). The exclusion criteria were: (1) Animal studies, family-based studies, the studies reporting parents who have children with NSCL/P as cases and/or parents without NSCL/P children as controls, erratum studies, case reports, review articles, systematic reviews, and conference papers and (2) studies without having the required data to calculate the genotype distributions.The inclusion criteria were: (1) Human and unrelated case-control studies, (2) studies including NSCL/P patients with no other systematic disease and healthy controls, (3) studies reporting rs13041247 and/or rs11696257 p-value of the Hardy–Weinberg equilibrium (HWE) in controls, and the quality score were extracted from the articles included in this meta-analysis. Data extraction was independently performed by two authors (M.S and P.L.-J).The first authors’ name, publication year, ethnicity, controls’ source, number of NSCL/P patients and controls based on each genotype and related polymorphisms, genotyping method, The Newcastle–Ottawa scale (NOS) was applied for the quality assessment of articles with a maximum score of nine . The quaThe qualitative synthesis, which includes assessment of bias and heterogeneity, describes the characteristics and findings of retrieved studies based on the extracted data. The estimated effects with CIs for each study analyzed in the meta-analysis were reported based on five genetic models  to have MAFB polymorphisms and risk of NSCL/P. The pooled OR significance was shown by the Z test, with a p-value < 0.05. The Chi2 test, the Tau2, and the I2 statistic were used to estimate heterogeneity between the studies and p-value < 0.1 (I2 > 50%) presented statistically significant heterogeneity. If there was heterogeneity, the random-effects model was conducted to allow for probable differences in the results of studies (DerSimonian and Laird method) [The 95% CI and crude OR for each study were obtained by Review Manager 5.3 (RevMan 5.3) to present the association between the  method) . Otherwi method) . In addiMAFB polymorphism. We used the Comprehensive Meta-Analysis version 2.0 (CMA 2.0) software  to perform the funnel plot analysis [p < 0.05 (two-tailed) was considered as significant existence of publication bias). In addition, cumulative analysis was done after excluding one study to assess the stability of the results. If it did not change the pooled data, the stability or consistency of the analyses would be confirmed. The meta-regression as a possible source of heterogeneity was used with the p-value and regression coefficient (r) to assess the strength of the association among the study period, the sample size, and p-value of the HWE in controls with the risk of the rs13041247 polymorphism.The subgroup analysis was done based on ethnicity, controls’ source, and genotyping method for rs13041247 analysis  using boOne hundred and two articles were retrieved from the databases. After removing the duplicates, 35 articles were screened . Of 35 aMAFB polymorphism involving 3082 NSCL/P patients and 4104 controls. Moreover, three studies reported MAFB rs11696257 polymorphism involving 845 NSCL/P patients and 927 controls. The p-value for HWE in controls of all studies was more than 0.05, and several genotyping methods were employed in the studies. In one of the studies [Some basic features of eight articles reviewed in this meta-analysis are presented in  studies , two casMAFB polymorphism based on five genetic models. The results identified that the pooled ORs of C vs. T, CC vs. TT, TC vs. TT, TC + CC vs. TT, and CC vs. TT + TC were 0.88 ), 0.68 ), 0.81 ), 0.79 ), and 0.82 ), respectively. Therefore, the risk of NSCL/P could be related to the CC and TC genotypes of rs13041247 MAFB polymorphism.MAFB polymorphism based on five genetic models. The pooled ORs in the models of a vs. A, aa vs. AA, Aa vs. AA, Aa + aa vs. AA, and aa vs. AA + Aa were 0.90 ), 0.70 ), 0.95 ), 0.93 ), and 0.73 ), respectively. Therefore, the risk of NSCL/P could not be related to rs11696257 MAFB polymorphism.MAFB polymorphism are shown in The results of subgroup analysis for the risk of NSCL/P related to rs13041247 The total score of quality of all articles included in this meta-analysis is shown in p-values of previous results.Sensitivity analyses  on previous analyses did not change the pooled results. Therefore, previous results had optimal stability. In addition, leave-out-one-study analysis on the quality of the studies deleting one study with low quality  showed tp-value of the HWE were not the reasons for heterogeneity.Meta-regression analysis was performed for finding potential and possible sources of heterogeneity between rs13041247 polymorphism and susceptibility to NSCL/P . The resMAFB polymorphisms. The Begg’s and Egger’s tests did not reveal any publication bias across the studies in each analysis (p > 0.05).MAFB and the risk of NSCL/P. The rs11696257 polymorphism was not associated with the risk of NSCL/P, but the CC and TC genotypes rs13041247 polymorphism was associated with a significant decline in the risk of NSCL/P and had a protective role against NSCL/P with a heterogeneity of 80% for CC genotype and 51% for TC genotype. This significant decline in the risk of NSCL/P was observed in population-based studies with lower heterogeneities for the C allele and CC genotype  or even lack of heterogeneity the TC genotype.NSCL/P is a complex congenital anomaly to present both clinical and genetic heterogeneities . This mep-value of the HWE in controls), and also, there was no regular pattern when we examined the forest plots ordered by these variables. The expression of The MAFB mRNA and protein in the craniofacial neuroectoderm and the mesoderm are derived from the neural crest between embryonic days 13.5 and 14.5 [MAFB has been shown as a candidate gene in NSCL/P pathogenesis according to genome-wide association study (GWAS) findings [MAFB on 20q12 were not associated with NSCL/P achieved genome-wide significance [MAFB and other genes in both European and Asian populations with NSCL/P [Out of 10 articles ,26,27,28and 14.5 . This inand 14.5 . MAFB hafindings . One stufindings  on 298 cfindings . Anotherh NSCL/P .MAFB is scarce in the literature, especially in human models. Some studies [MAFB gene promotes the differentiation of some cells, such as phagocytic, β-islet, bone, and glomerular epithelial cells. In addition, this gene has been related to serum lipid concentrations, and also, the ischemic heart disease and stroke risk, which means that MAFB gene interacts with higher BMI, hypertension, and diabetes [MAFB gene. Thus, the gene–gene and gene–environment interactions should be considered when assessing the association of polymorphisms of this gene with the risk of NSCL/P and other diseases.The information regarding  studies ,34,35 rediabetes . TherefoThis meta-analysis had several limitations such as (1) small number of studies available on this topic, (2) various genotyping methods, (3) high heterogeneity across the studies, (4) the Begg’s and Egger’s tests have low power, so it is difficult to exclude publication bias, and (5) the multiplicity of testing in the methods. However, the high quality of most studies and stability of the results were the strengths of this meta-analysis.MAFB. The CC and TC genotypes are significantly associated with the decreased risk of NSCL/P. In addition, a significant association was noted between rs13041247 polymorphism and a reduction in risk of NSCL/P in population-based studies. Well-designed studies are required to assess the interaction of MAFB gene and other genes with environmental factors in different ethnic groups.The results of the present meta-analysis show that the risk of NSCL/P is related to rs13041247 polymorphism, not rs11696257 polymorphism of


WNT3 gene polymorphisms and non-syndromic cleft lip (CL) with or without cleft palate (NSCL/P) predisposition. Methods: A comprehensive literature search was performed in six online databases including PubMed, Embase, ISI Web of Science, CENTRAL, CNKI, and Wanfang from inception up to June 2018 without language restriction. Pooled odds ratios (ORs) and corresponding 95% confidence intervals (95%CIs) were calculated under allele model of inheritance to indicate the association between WNT3 polymorphisms and NSCL/P. Risk of bias was assessed through the Newcastle–Ottawa scale (NOS). Predetermined stratified and sensitivity analyses were performed using the RevMan 5.3 software, publication bias were evaluated by Egger’s and Begg’s tests. Results: Seven case–control studies comprising 1617 NSCL/P patients and 2143 healthy controls were identified and included in the present study, a total of eight loci were investigated in the present study: rs3809857 was significantly associated with NSCL/P vulnerability , a significant association between rs9890413 polymorphism and NSCL/P susceptibility  was detected as well. Since only few studies reported detailed data about the association between rs142167, rs7207916, rs199498, rs111769, rs12452064, rs11653738, and NSCL/P risk, these results were not combined using meta-analysis. Conclusion: Based on the findings of our current study, the rs3809857 and rs9890413 polymorphisms of WNT3 appeared to be associated with NSCL/P. Limited evidence is found to support the association between other WNT3 polymorphisms and risk of NSCL/P.Objective: This meta-analysis was conducted with the aim of investigating the association between  Non-syndromic cleft lip (CL) with or without cleft palate (NSCL/P) is one of the most frequent congenital malformation amongst live births worldwide. The global prevalence of NSCL/P ranging from 1:500 to 1:2500 live births, with considerable variability across geographic origin, ethnic and racial background, as well as socioeconomic status . AlthougIRF6 [TGFA [MTHFD1 [MTHFR [BMP4 [As a complex disorder, current knowledge about the etiopathogenesis of NSCL/P remains largely undetermined. Environmental variations, genetic influences and interactions of these factors all drive the occurrence of NSCL/P. Researchers have reported that exposure to environmental factors like drinking, tobacco smoking, medicinal drugs, viral infection, and lack of vitamins during early pregnancy appear to increase the risk of having NSCL/P offspring ,6. ApartIRF6 , TGFA [9 [MTHFD1 , MTHFR [1 [MTHFR , BMP4 [1FR [BMP4 , and so WNT gene family consists of a range of members such as WNT3, WNT3A, WNT5A, WNT8A, WNT11, and so forth [WNT genes constitute a cluster of conserved secreted glycoproteins that participant in developmental and cell-biological processes, thus playing a fundamental role in craniofacial embryogenesis [WNT genes and their downstream targets have been implicated in human NSCL/P [WNT family, WNT3 gene is the most extensively studied candidate gene for NSCL/P.The so forth . The proogenesis ,15. Mutan NSCL/P . Of the WNT3 gene polymorphisms and NSCL/P vulnerability, and found some locus polymorphisms significantly associated with NSCL/P. Afterward, several replicate studies were performed across different races with inconsistent outcomes [WNT3 polymorphisms and NSCL/P predisposition.Early in 2008, Chiquet et al.  initialloutcomes . RespectThe current systematic review and meta-analysis was performed in accordance with PRISMA  guidelines . An unpuThe computerized literature search of the present study was systematically conducted using six online electronic databases including PubMed, Embase, ISI Web of Science, CENTRAL, CNKI, and Wanfang. All the relevant literature was published previous to June 2018. A combination of Medical Subject Headings (MeSH) alongside free terms was utilized to hunt all the potentially eligible publications without any language restriction. For English databases, we employed the following search string: , and , and Wnt3. For Chinese academic databanks, we adopted the key words to identify related Chinese literature, the literature search strategy was as followed: ‘Chun E Lie’, ‘WNT3’, and ‘Duo Tai Xing’. The reference lists of relevant reviews and full-text articles were manually examined as well to look for additional possible studies.WNT3 polymorphisms and NSCL/P susceptibility that have been published; (iii) control subjects should be defined as healthy subjects without history of cleft and other major congenital anomalies; (iv) available odds ratio (OR) and 95% confidence interval (95%CI) under allelic comparison of individual loci; (v) observational studies (case–control or cohort studies) on humans.The PICOS principle was followed to establish the inclusion criteria for our current study. Studies satisfying the following criteria were included for review: (i) NSCL/P should be diagnosed on the basis of clinical examination; (ii) studies on the relationship between Accordingly, studies would be excluded if they met the following criteria: (i) animal studies, reviews, case reports, conference abstracts, as well as editorials; (ii) data that overlapped previous publications. If duplicated studies reporting overlapping data were identified, the most comprehensive one was included in the meta-analysis.The methodological quality of eligible studies was evaluated separately by the first two investigators according to the Newcastle–Ottawa scale (NOS) for observational studies . A ‘starIn compliance with the predefined criteria, the following information was meticulously extracted independently by two reviewers (B.-q.W. and S.-t.G.) from all qualified articles: (i) surname of the first author; (ii) year of publication; (iii) variant locus; (iv) country where the study was performed; (v) ethnicity of enrolled subjects; (vi) numbers of case and control subjects; (vii) results of Hardy–Weinberg equilibrium (HWE) test; (viii) OR and 95%CI under allelic comparison of individual locus; (ix) major allele distribution of case and control participants. Once encountering discrepancy during this process, two authors re-inspected the article together and reached an agreement after discussion.WNT3 polymorphisms and NSCL/P susceptibility was evaluated through merging ORs and corresponding 95%CIs of individual studies. The between-study heterogeneity was assessed using the Q-statistical test and 2I test [P<0.1, 2I > 50%) or absence of heterogeneity , respectively [The strength of association between  I2 test . The ranectively ,29. The ectively . Forest in silico tools were applied to predict the function of WNT3 polymorphisms: HaploReg 4.1 [http://pubs.broadinstitute.org/mammals/haploreg/haploreg.php) and GTEx Project [https://gtexportal.org/home/).Two oReg 4.1  , the fixed-effect model was employed for calculating the association between rs3809857 and NSCL/P. The finding indicated that rs3809857 polymorphism was significantly associated with NSCL/P vulnerability , the random-effect model was applied for data combination. The result suggested a null association between rs9890413 polymorphism and NSCL/P vulnerability , while the study by Xin et al. [P=0.29).Two studies reported rs142167 polymorphism and NSCL/P susceptibility with similar sample size but conflicting outcomes. Menezes et al.  found rsn et al.  found a P=0.29).Only one study investigated rs7207916 polymorphism and NSCL/P susceptibility . No signP=0.01), yet rs111769 polymorphism tended to reduce NSCL/P risk .Menezes et al.  reportedP=0.86).Mostowska et al.  studied P=0.0064).Nikopensius et al.  investig2I = 4%, P=0.37). Therefore, this study was deemed as the source of heterogeneity across included studies, as a result we excluded this study from the meta-analysis and recalculated the OR and 95%CI for the association of rs9890413 and NSCL/P. The outcome became stable when performing sensitivity analysis again, and indicated a significant association between rs9890413 polymorphism and NSCL/P susceptibility  and Begg’s test  for rs3809857 and rs9890413 also suggested no statistically significant publication bias. Subgroup analyses by type of NSCL/P and ethnicity were not conducted in the present study because of limited number of included studies.The leave-one-out sensitivity analyses were employed to test the robustness and reliability of the outcomes. For rs9890413, when the study by Lu et al.  was remoP=1.8e-7; P=3.3e-7; P=4.6e-7; P=0.0000024; P=0.000053; P=0.000085; The GTEx database was used to study the relationship between genetic variations and gene expressions in human tissue. We found that the T allele of rs3809857 is associated with lower expression in testis . In comparison with other included studies, Lu et al. reported a much higher A allele frequency in both NSCL/P group (Lu: 92.6% compared with mean: 81.0%) and control groups (Lu: 96.9% compared with mean: 79.0%). As a result, study by Lu et al. [Majority of included studies presented insignificant correlation between rs3809857 and rs9890413 polymorphisms and NSCL/P vulnerability. Nevertheless, the pooled data indicated significant outcomes, which suggested restricted statistical power of each individual study. For rs9890413, when the study by Lu et al. was removed, the overall effect changed and the heterogeneity became insignificant , G allele of rs3809857 is associated with higher WNT3 expression in human testis (2 ≥ 0.8. Notably, rs17692129 and rs35937770 are both polymorphisms within intron of NSF gene. As far as we know, NSF is involved in intracellular protein transport, but the impact of NSF polymorphisms on NSCL/P has not yet been confirmed. Taken together, both rs3809857 and rs9890413 have regulatory motifs changed, and both the variants are associated with altered expression of WNT3 in human organs, which could be one of the reasons why these polymorphisms confer risk of NSCL/P. To test these hypotheses, more functional experiments are still needed.As the most abundant type of genetic variant, SNPs contribute substantially to phenotypic differences amongst individuals. The past two decades have witnessed great advance in the mapping of loci for NSCL/P. Regarding research on t et al.  first inn testis A, pancren testis B, and thn testis C, while n testis D, tibialn testis E, and lun testis F. Furthen testis A and rs9n testis B. AccordWNT3 polymorphisms and the risk of NSCL/P vulnerability. To improve the interpretation of the results, some limitations of the present study should be considered. First of all, in terms of the sample size, individual studies used hundreds but not thousands of participants, thus lacking adequately statistic power to guarantee the association. Second, we only investigated the role of eight loci polymorphisms separately. However, like most complex disorders, NSCL/P is caused by multiple loci polymorphisms and a string of genes with synergetic effects. Interactions of these loci and genes might conceal or magnify the actual function of a single locus polymorphism. Third, only five articles in English and two in Chinese from six databases were retrieved for the meta-analysis, potential relevant articles published in other languages might have been skipped, which might incur inclusion criteria bias. Fourth, due to lack of available data, we only conducted meta-analysis under allele model. At last, as all the component studies were hospital-based, the selection bias of participants might be introduced, hence limiting the extrapolation to general population of the current findings.As far as we have known, the present study is the first comprehensive meta-analysis of the potential association between the WNT3 polymorphisms and NSCL/P risk, based on the results of our meta-analysis we did find some genetic variants in WNT3 gene that possibly contribute to the risk of NSCL/P. The present study suggested that G allele of rs3809857and A allele of rs9890413 polymorphisms in WNT3 were significantly associated with increased NSCL/P vulnerability. Both rs3809857 and rs9890413 have regulatory motifs changed, and both the variants are associated with altered expression of WNT3 in human organs, which could be one of the reasons why these polymorphisms confer risk of NSCL/P. Insufficient evidence is found to support the association between other WNT3 polymorphism and risk of NSCL/P. Concerning limitations of the current study, our findings need further confirmation by well-designed and population-based investigations with larger sample size amongst more ethnicities.More evidence is needed to uncover the association between