Egger’s linear regression test did not show a potential publication bias , which was also confirmed by the symmetry of the funnel plot. Moreover, no high statistical heterogeneity  was found. The results of this meta-analysis showed a potential protective effect of GT consumption on BC, especially for BC recurrence.Breast cancer (BC) is the most common cancer in women and several factors are involved in its onset. Green tea (GT) has been shown to have potential beneficial effects on different types of cancer. The aim of this review was to evaluate the association between GT regular consumption and risk of BC in women. The risk of BC recurrence and risk of BC in relation to menopausal status were also evaluated. A literature search of PubMed, Scopus, and Web of Science was conducted. Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines were followed to perform the systematic review and meta-analysis. Full texts were downloaded for 40 studies; however, only 13 records were included in the meta-analysis. Eight were cohort studies and five were case-control studies. The pooled sample consisted of 163,810 people. An inverse statistically significant relationship between GT and BC risk, with an Odds Ratio (OR) = 0.85 ((95% CI = 0.80–0.92),  Camellia sinensis (C. sinensis). The fruit of C. sinensis is three-celled and green in color, and the seed is a brown, semi-globose nut  OR “Camellia sinensis”[MeSH] OR “Tea”[MeSH] OR “green tea”[Title/Abstract] OR “Polyphenols”[MeSH] OR “epigallocatechin gallate”[Supplementary Concept];-breast cancer risk: ((“breast neoplasms”[MeSH Terms] OR breast cancer[Text Word] OR breast cancer[Title/Abstract] OR ((“breast”[MeSH Terms] OR “breast”[All Fields]) AND tumor[Title/Abstract];-study design: “Case-Control Studies”[MeSH] OR “Cohort Studies”[MeSH] OR “Retrospective Studies”[MeSH] OR “Prospective Studies”[MeSH].For searches on Scopus, the following search terms were used:-green tea consumption: “green tea” OR “Camellia sinensis” OR Tea OR epigallocatechin gallate OR Polyphenols-breast cancer risk: breast neoplasm OR breast cancer OR breast tumor-study design: “Case-Control Studies” OR “Cohort Studies” OR “Retrospective Studies” OR “Prospective Studies”.Articles included in the current meta-analysis met the following criteria: studies performed on humans, research focused on green tea consumption, full text available, epidemiologic studies, only articles in English. Inversely, exclusion criteria were: studies with different outcomes, data not reported as risk  or articles without original data, and animal model trials. As the aim of this review was not to evaluate the effectiveness of an intervention in a population, randomized controlled trials (RCTs) were not included . FurtherTo validate the inclusion of the retrieved studies, two reviewers (V.G. and A.A.), independently and blinded, screened titles and abstracts. Possible disagreements were resolved through discussion or third reviewer consultation (D.N.). The reviewers read the full texts of the selected articles and double-checked the list of references of the selected studies in order to detect any other potentially relevant papers.Data were collected independently by two authors (V.G. and A.A.) and reported in a pre-determined spreadsheet. Extracted data contained both qualitative and quantitative information. In particular, qualitative data included the name of the first author, year of publication, type of study, and country where the study was conducted. Quantitative study design characteristics, namely data on enrolment, exposure, tools used to estimate food intake , validated or not, with the FFQ either self-administered or by interview), and information on the outcomes, were also collected. Furthermore, participant characteristics such as sample size, age range or mean age, and health status were also collected.i) justification given for the cohort ; (ii) appropriate inclusion and exclusion criteria; (iii) histological diagnosis of breast cancer; (iv) validated tool to assess green tea consumption; (v) adjustments were made for age, sex, body mass index, smoking status, and parity; and (vi) any other adjustments .In order to perform a sensitivity analysis, the quality of the included studies was also evaluated by two authors (A.A. and M.A.) using the scoring system created on the basis of Meta-analysis of Observational Studies in Epidemiology (MOOSE) , Qualityp < 0.05 was considered statistically significant. The statistical heterogeneity among studies was assessed by the Chi2 test and I2 statistic  [The software ProMeta 3  was used to run the meta-analysis. The effect size (ES) was estimated by odds ratio (OR) reported with its 95% confidence interval (CI). For this study, geneous) . To calcThe meta-analysis was conducted considering the risk of breast cancer among participants exposed to the highest versus the lowest consumption of green tea, as reporting of the frequency, dose, and unit  of tea consumption varied among the original studies.In order to corroborate the strength of the results, an additional subgroup analysis was performed, taking into account quality score, study design, risk of BC, and recurrence.From PubMed/Medline, 47 articles were retrieved, whilst 56 articles were found in Scopus and 90 in Web of Science. After bibliography list screening, one more article was added, giving a total of 194 potentially relevant articles. However, 39 studies were duplicates and immediately eliminated. After further screening, based on evaluation of title and abstract, another 115 records were excluded, as reported in p = 0.000)) , also confirmed by the symmetry of the funnel plot  was found. All the studies included in the analysis are tabulated in  0.000)) a. The renel plot b. No higp = 0.273)) and with low potential publication bias . However, when the two studies that enrolled women with BC were excluded, the results did not change significantly ES = 0.97 ((95% CI = 0.86–1.11), p = 0.273)) . Inversely, after selective inclusion of only case-control studies, the pooled sample size consisted of 12,324 participants with ES = 0.81([(95% CI = 0.74–0.89), p = 0.000)) and with no potential publication bias , even though a statistical heterogeneity was found ; probably due to a low number of included studies. Data are shown in Any potential change in the results was analyzed, firstly performing the analysis according to the original studies design. When only cohort studies were included in the meta-analysis, the selection ranged from 472 to 67,422 participants followed up for 6 years on average. All the studies recruited healthy participants except for two papers that selected women with BC diagnosis. Considering all the cohort studies, the pooled sample size consisted of 151,486 participants with ES = 0.94 ((95% CI = 0.83–1.05), p-value = 0.000) based on a total of 13,956 participants, with participant numbers in individual studies ranging from 472 to 6928 participants . The funnel plot .Seven studies focused on the risk of BC recurrence ,69,71,72icipants a. A slignel plot b shows np-value = 0.684). However, no statistical heterogeneity  was found (data not shown). Moreover, only three studies reported the stage of BC. In particular, Inoue et al. 2001 and Iwasaki et al. 2014 [p = 0.926) based on 1898, participants with no statistical heterogeneity .A supplementary analysis of the studies focusing on the risk of new BC diagnosis was conducted. In this case, none of the included studies reported a statistically significant ES, which was also confirmed by the pooled ES of the current meta-analysis. Indeed, the combined ES was 0.97 ((95% CI = 0.86–1.11), al. 2014  only incp-value = 0.000) based on a total of 112,300 participants, with the number of participants for individual studies ranging from 738 to 67,422 participants . The funnel plot .For a more comprehensive evaluation, a meta-analysis of studies with QS ≥ 5 was performed. In this case, ES was 0.73 ((95% CI = 0.63–0.85), icipants a. A slignel plot b shows pp-value = 0.783) based on a total of 148,511 participants, with the number of participants for individual studies ranging from 427 to 67,422 participants. No statistical heterogeneity was found , and the lack of potential publication bias was confirmed by Egger’s linear regression test .In order to find a potential beneficial amount of green tea, another meta-analysis of the studies reporting 5 cups of tea/day was performed. In this case, ES was 0.97 ((95% CI = 0.81–1.18), A meta-analysis looking at menopausal status was also performed . p-value = 0.035) based on a total of 1729 participants, with participant numbers for individual studies ranging from 79 to 1302 participants , with the lack of potential publication bias confirmed by Egger’s linear regression test   based on a total of 1486 participants, with participant numbers for individual studies ranging from 70 to 799 participants. No statistical heterogeneity was found , with potential publication bias confirmed by Egger’s linear regression test .The risk of BC in women before and after menopause was compared. In the analysis, green tea intake showed a statistically significant protective role in pre-menopausal women. In this case, ES was 0.88 ((95% CI = 0.78–0.99), icipants a. No sta= 0.887) b. InversThe potential beneficial effect of green tea could be related to its high amount of natural phytochemicals, which are preserved because of the absence of fermentation during the production process. Among green tea phytochemicals, polyphenols are the most represented. Several experimental studies have shown that long-term exposure to polyphenols may reduce chronic inflammation and oxidative stress and inhibit the growth, reproduction, and diffusion of cancer cells ,74. ThesThis meta-analysis showed the potential effect of green tea consumption in the reduction of breast cancer risk (15%). Furthermore, because the CI shows the interval within which a true mean is likely to be found, the narrow CIs observed in this systematic review may suggest the precision of ES results. In a sensitivity analysis of case-control studies and those with QS ≥ 5, a higher protective effect of green tea was found. In particular, a 19% reduction for case-control studies and a 27% reduction for studies with QS ≥ 5 were found. The majority of cohort studies included in this meta-analysis found a significant reduction of BC recurrence, although data was not confirmed in the case-control studies. The presented results highlight the important role of green tea in tertiary prevention rather than in primary prevention. Actually, green tea appears to reduce BC incidence in only 3% of the cases and not in a statistically significant manner. Inversely, the 19% reduction of BC recurrence appears to be associated with green tea drinking. However, green tea consumption was assessed in all the studies through questionnaire or interview, though only five studies used validated tools. Even though the food frequency questionnaires are cheaper and more manageable instruments to assess dietary intake, they cannot be considered devoid of potential bias. It was not possible to standardize the appropriate consumption value because green tea consumption was recorded in different units in the original studies. Indeed, tea consumption was reported in grams or cups. Furthermore, when cup unit was used as a reference, the amount of tea per cup differed, ranging from 100 mL to 350 mL. Due to this limitation, it was not possible to perform a dose–response analysis. In the absence of clear evidence answering questions about the right amount and frequency of consumption, it is not possible to estimate the amount of tea that may provide preventive effects. Indeed, a comparison between the highest intake of green tea, as reported in the original papers, with no consumption or lowest intake was performed. Nevertheless, a subgroup analysis that only included studies with green tea intake higher than 5 cups/day was carried out. Moreover, in order to minimize potential bias, adjusted data were collected from the original studies. In particular, attention was focused on adjustment for age, Body Mass Index (BMI), parity, history of BC, and smoking habit. Nonetheless, several studies also adjusted for other potential confounding factors such as age of menarche, age of first pregnancy, number of births, and menopausal status. This meta-analysis shows similar results previously obtained in another meta-analysis performed in 2010 . HoweverSome limitations have to be taken into account when reading this meta-analysis. First, the majority of the studies were conducted in Asia, mostly China and Japan, which represent the largest green tea production and consumption areas. Only one study was conducted in the USA, and no studies were conducted in Europe. This highlights the importance of performing similar studies in other ethnic populations in order to underline any potential differences. This is particularly true in determining whether other antioxidants consumed in the diet can play a confounding role. Another limitation is that self-reported green tea intake by FFQ was measured at baseline only, and the duration of tea consumption is not known. This can result in errors during both reporting and classification. Moreover, the different amounts of green tea consumed as higher intake in the original studies could have been defined in post hoc analysis. This is likely the first meta-analysis also assessing green tea intake, the risk of BC in post-menopausal women, and BC stage. Another significant strength of the present study is that the risk for BC and recurrence were considered both together and separately. However, due to the absence of information in the original studies, it was not possible to evaluate the effect of green tea based on BC stage at diagnosis. All the above-mentioned limitations can explain the different results between studies and the absence of statistical significance when looking at the different states of menopause or the preferable amount of intake. Future studies need to focus on these particular aspects. Moreover, studies with higher numbers of participants need to be performed. This meta-analysis has some important strengths, such as the large number of participants , with participants ranging from 20–87 years of age. Another strength is represented by the sensitivity analysis that was conducted in population subgroups. Furthermore, results in population subgroups are important for public health decision-makers to help them implement effective evidence-based strategies for selected population groups , thereby enhancing the quality and efficiency of interventions . According to the present results, the highest protective effect of green tea was found in BC recurrence prevention. This is in line with previous studies, where the authors demonstrated an anti-neoplastic effect of green tea through a reduction in viability and proliferation of BC cells ,78,79. IFurther epidemiological studies are needed to confirm these results and to quantify the impact of behavior change in this disease. Moreover, the majority of the included studies performed a satisfactory follow-up (a mean of approximately 10 years). A statistically significant inverse association between green tea and BC was found. This finding is in line with previous publications. However, the results are not always concordant, especially when looking at cohort studies or selecting only those studies analyzing the diagnosis of BC recurrence. However, a beneficial effect of green tea cannot be excluded, in particular because when only case-control studies were analyzed, a statistically significant protective effect was observed. Indeed, the importance of case-control studies in defining the causal relationship between exposure and event is well known. Therefore, further studies are required to confirm these data before any public recommendations can be made. In particular, it is important to conduct studies with greater numbers of people, in different countries, and with a serial evaluation of lifetime green tea consumption.Moreover, as tea is one of the most popular beverages in the world and may possess many chemo-preventive qualities, a better understanding of the mechanisms might improve the utilization of green tea in breast cancer primary and recurrence prevention. At the clinical level, results from this systematic review and meta-analysis may be useful to design clinical trials aimed at ascertaining the potential role of green tea as an adjuvant in breast cancer therapy and pave the way to novel preventive strategies for breast cancer.


MMP-9 polymorphisms  and asthma susceptibility are inconclusive. To derive a more precise estimation of this association, a meta-analysis was performed. A literature search was conducted in PubMed, Web of Science, EMBASE, Wanfang, and China National Knowledge Infrastructure (CNKI) databases to identify eligible studies. The pooled odds ratios (ORs) and corresponding 95% confidence intervals (CIs) were used to calculate the strength of association. Sensitivity analysis was performed to evaluate the influence of individual studies on the overall effect estimates, and funnel plots and Egger's test were inspected for indication of publication bias. Seven studies with 1592 asthma patients and 1987 controls were finally identified. Overall, we found no significant association between −1562 C > T, rs3918242 polymorphism, and asthma susceptibility in any of the genetic model comparisons. After categorizing studies into different subgroups on the basis of ethnicity and age, there is still no significant association. For the Gln279Arg, rs17576 polymorphism, there seems to be a significant association in the allelic genetic model in regard to the P value =0.044); however, the value of lower 95% CI is 1.0. For the Arg668Gln, rs17577 polymorphism, a high significant association was observed in the dominant model comparison =0), recessive model comparison =0.011), homozygote genotype comparison =0.004), and allelic genetic model =0). Sensitivity analysis demonstrated the stability of our results, and publication bias was not evident. The present meta-analysis suggests that MMP-9 Arg668Gln, rs17577 polymorphism may be the risk factor for asthma susceptibility.Published data on the association between  In thmponents .MMP-9, one of the MMP family proteins, is mainly produced by macrophages and neutrophils but also by epithelial cells, mast cells, fibroblasts, and smooth myocytes, which are key mediators in the airway wall remodeling and metabolism of the extracellular matrix [MMP-9 to its natural inhibitor (TIMP-1) in bronchoalveolar lavage (BAL) fluid was also higher in children with symptomatic asthma, as compared to that of healthy controls [MMP-9 in mice leads to enhanced allergen-induced airway inflammation and increases the numbers of eosinophils and levels of cytokines such as interleukin (IL)-4 and IL-13 [MMP-9 plays an important role in asthma pathogenesis.r matrix –7. Its ir matrix , 9. The controls . In addind IL-13 –13. ThesMMP-9 gene is located on chromosome 20q11.2–q13.1, a position which has been shown to be associated with bronchial hyperresponsiveness and specific sensitization [MMP-9 expression and function [MMP-9 gene polymorphisms and susceptibility of asthma in different populations [MMP-9 gene, the association of Gln279Arg, rs17576 and Arg668Gln, rs17577 polymorphisms with the susceptibility of asthma was also assessed [MMP-9 gene and asthma susceptibility. However, these results were inconclusive and inconsistent. Therefore, we performed a meta-analysis of all eligible studies to obtain more precise estimation of the association of MMP-9 gene polymorphisms including three SNPs  with asthma susceptibility.The tization , 15. Up function –18. Therulations –26. Mostulations –21,23,25assessed ,26. TherMMP-9 or Gelatinase B in combination with polymorphism or mutation or variant. Searching was performed in duplicate by two independent reviewers.Publications were searched using the Pubmed, EMBASE, Web of Science, Chinese National Knowledge Infrastructure (CNKI), and Wanfang databases . The search strategy utilized in our study was as follows: asthma or asthmatic and matrix metalloproteinase 9 or The inclusion criteria of our study were as follows: (1) Any human studies that estimated the prevalence of matrix metalloproteinase 9 polymorphisms and asthma risk were included, which are published in English or Chinese. (2) The genotype distributions or allele frequency of each study should be available for estimating an odds ratio (OR) with 95% confidence interval (CI). (3) There were sufficient results for extraction of data, that is, number of subjects for each genotype in asthma and control groups. Where eligible papers had insufficient information, we contacted authors by e-mail for additional information. Studies were excluded from our meta-analysis if their authors did not provide us with related data.The basic information extracted for each study was as follows: name of first author, publication year, country and ethnicity of case control, age of case, sample size, and genotype frequencies in cases and controls. Data were extracted independently and in duplicate by two reviewers who used a standardized data extraction form. Any disagreement was adjudicated by a third author.Newcastle-Ottawa Scale (NOS) was used to assess the quality of the included studies. Item assessed include selection, comparability of case/controls, exposure/outcome, age and gender. The quality scores ranged from 0 to 9. We divided NOS score into three levels .P < 0.05. The OR with 95% CI was used to assess the strength of the association between MMP-9 polymorphism and asthma risk. The pooled OR for MMP-9 polymorphisms and asthma risk was performed for dominant genetic model (AA + Aa vs. aa), recessive genetic model (AA vs. aa + Aa), homozygote genetic model (AA vs. aa), and allele genetic model (A vs. a), respectively. In the current study, the aa genotype was a wild-type, while the AA genotype was a mutant. The heterogeneity was assessed by using Q-test and I2 test. A P value > 0.10 of Q-test and I2 < 50% indicates a lack of heterogeneity among the studies, then the fixed-effect model was used. Otherwise, the random effect model was used. Subgroup analysis was performed by ethnicity and case age to further explore ethnicity-specific and age-specific effects. Sensitivity analysis was conducted by sequentially excluding one study at a time to examine the effect of each study on the combined result. The funnel plot and Egger's test was used to assess the potential publication bias. All statistical analyses of this meta-analysis were performed using the STATA 11.0 software .Hardy–Weinberg equilibrium (HWE) was assessed for each study by use of Pearson's chi-square test in control groups, and significance was set at MMP-9 variants and asthma or not human studies. The remaining 15 articles were then assessed for inclusion. Of these, 8 articles were excluded because 3 articles are conference abstracts, 3 articles lacked a case-control design, and 2 articles lacked detailed genotypes. Eventually, we identified 7 case-control publications, including 1592 asthma patients and 1987 controls, to evaluate the association of MMP-9 polymorphisms  with asthma susceptibility [P > 0.05). These characteristics of each eligible study are shown in The flow diagram in tibility . There aMMP-9 −1562 C > T, rs3918242; Gln279Arg, rs17576; and Arg668Gln, rs17577 polymorphisms and asthma susceptibility are provided in Figures P value =0.044); however, the value of lower 95% CI is 1.0 =0), recessive model comparison =0.011), homozygote genotype comparison =0.004), and allelic genetic model =0) =0) .For −1562 C > T, rs3918242 polymorphism, sensitivity analysis was conducted by sequentially excluding individual studies to estimate the stability of the results. After sequentially excluding each study, statistically similar results were found . PotentiMMP-9 gene polymorphisms and susceptibility of asthma. Our results indicated that there was no significant effect of MMP-9 −1562 C > T, rs3918242 and Gln279Arg, rs17576 polymorphisms on asthma susceptibility in overall analyses and subgroup analyses. However, for Arg668Gln, rs17577 polymorphism, a significant association with asthma susceptibility was observed in overall analyses.To the best of our knowledge, this is the first meta-analysis to assess the association between MMP-9 level is not only observed in asthma, but also existed in others inflammatory disease, such as acute respiratory tract diseases and chronic obstructive pulmonary disease (COPD) [MMP-9 or other genes involving in the inflammatory response which play more fundamental roles in asthma. Indeed, one study showed a strong linkage disequilibrium between the −1562 C > T, rs3918242 polymorphism and 2127 G > T, rs2274755 polymorphism in the fourth intron, increasing the transcriptional level of MMP-9 and involving in the development of asthma [The −1562 C > T, rs3918242 polymorphism in the promoter sequence is the most frequently studied and best-recognized polymorphism, which is important for gene transcription rate and protein level . Howevere (COPD) . So the MMP-9. The Arg668Gln replacement may result in increased MMP-9 resistance to its natural inhibitors and higher in vivo activity of this enzyme [MMP-9 in children with asthma [P value; however, the value of lower 95% CI is 1.0. Therefore, more well-designed and high-quality studies with a larger sample size among the different populations should be conducted to support this finding.Besides the −1562 C > T, rs3918242 polymorphism in the promoter region, we analyzed the other two polymorphisms in the coding region, namely, Gln279Arg, rs17576 and Arg668Gln, rs17577. Gln279Arg, rs17576 and Arg668Gln, rs17577 polymorphisms are single nucleic acid substitutions at positions 836 (A-G) and 2003 (G-A), corresponding to substitutions of glutamine for arginine at amino acid position 279 and arginine for glutamine at amino acid position 668. The Gln279Arg, rs17576 polymorphism is located close to the catalytic region of this enzyme. This change is supposed to modify the affinity of catalytic region of s enzyme . In addih asthma . These rSeveral limitations should be taken into account when interpreting our results. First, despite a comprehensive search, the number of studies that qualified for inclusion was modest, and the results might be exposed to interference factors such as random error. Second, only published data were included, leading to possible publication bias in this meta-analysis, although no statistically significant publication bias was identified. Third, even though the existing literature had acceptable quality, detailed information was not provided, for e.g., asthma definition varied among different articles, and this may be a confounding factor. In addition, participation rates for cases and controls were not reported in the majority of included studies, thus our meta-analysis was unable to explore the selection bias. Moreover, with limited information about environmental risk factors for asthma, we have no access to evaluate the gene-gene and gene-environmental interactions. Fourth, all the studies emphasize the fact that the differences in polymorphic variation depend on age, ethnicity, and gender, while ignoring the effects of epigenetic modulation.MMP-9 −1562 C > T, rs3918242 and Gln279Arg, rs17576 polymorphisms are not the independent risk factor for asthma susceptibility; however, the Arg668Gln, rs17577 polymorphism is the risk factor for asthma susceptibility.In conclusion, the current meta-analysis indicates that the


Objectives: TP53 is an important tumor suppressor gene to maintain genomic integrity, and its mutations increase the susceptibility to oral carcinoma. Previous published studies have reported the relation of TP53 codon 72 polymorphism with the risk of oral carcinoma, but the results remain controversial and inconclusive.Methods: We therefore utilized meta-analysis based on a comprehensive search in PubMed, EMBASE, and Google of Scholar databases up to August 19, 2017.Results: Total 3,525 cases and 3,712 controls from 21 case-control studies were selected. We found no significant association between TP53 codon 72 polymorphism and oral carcinoma susceptibility in all genetic contrast models, including subgroup analysis based on control source and ethnicity. Furthermore, TP53 codon 72 polymorphism was not significant associated with oral carcinoma susceptibility in tobacco or alcohol use, and HPV infection status. Our results were confirmed by sensitivity analysis and no publication bias was found.Conclusions: Taken together, our data indicate that TP53 codon 72 polymorphism is not associated with the susceptibility to oral carcinoma. Based on the GLOBOCAN2012 investigations, oral carcinoma is regarded as one of the most common causes of cancer related morbidity and mortality, contributing to 3.8% of all cancer cases and 3.6% of cancer related deaths ].PubMed, EMBASE, and Google of Scholar databases up to August 19, 2017 were searched with a combination of the keywords as follows: [ AND (tumor OR carcinoma OR cancer) AND (TP53 OR P53 OR Arg72Pro) AND  genotype distribution was in Hardy-Weinberg equilibrium (HWE). Major exclusion criteria were: (i) Reviews, conference abstracts, case reports; (ii) only-case study; (iii) genotype distribution was inconsistent with HWE; (iv) when duplicated studies published, only the study with the large sample size was included.Inclusion criteria were: (i) evaluated the association between tobacco and/or alcohol uses, TP53 codon 72 polymorphism, HPV infection, and susceptibility to oral carcinoma; (ii) case-control researches published in English or Chinese; (iii) definite histopathologic diagnosis or clearly reported the type; (iv) sufficient data to evaluate the ORs and 95%CI, and Data extraction was performed independently by two authors using a standardized form. Data such as: first author, country, ethnicity, year of publication, source of the controls, genotype distribution of cases and controls. Discrepancies were settled by discussion, with disagreements resolved by consensus.Q-test and I2 statistics were used to quantify statistical heterogeneity. The random-effect model was conducted if the heterogeneity was significant (P < 0.05)  with corresponding 95% credible interval (95%CI). P < 0.001) and three duplicated data publications were further excluded . Based on subgroup analysis, stratified by control source or ethnicity, we obtained similar results  was performed, and the association was still not significant in either tobacco users  or non-users . Similar results were found for subgroup analysis stratified by alcohol use or HPV infection status Figures .P < 0.05) was detected in some genetic comparisons, so random-effects model was adopted (DerSimonian and Laird, Between-study heterogeneity was examined and significant heterogeneity P < 0.0 was deteBegg's and Egger's test was utilized to examine the potential publication bias of the studies (Begg and Mazumdar, TP53 inactivation is a frequent event in cancer and involves point mutations and allelic loss (Baker et al., Although many case-control studies investigated the association of tobacco and/or alcohol uses, TP53 codon 72 polymorphism, and HPV infection with oral carcinoma susceptibility, the results were inconclusive. A past case-control studies failed to detect any significant association of TP53 codon 72 polymorphism with oral carcinoma susceptibility (Summersgill et al., Tobacco smoking is a well-known risk factor of cancer and could affect gene polymorphism in oral carcinoma (Ye et al., Regular alcohol consumption is associated with an increased risk for oral cancer. Such association is dose-dependent. Indeed, among individuals consuming 4–5 drinks daily, the risk for cancer of the oral cavity is 2–3-fold higher than among non-drinkers (Baan et al., HPV infection has been suggested as one of the contributing factors for oral carcinoma. It was suggested that the interaction of TP53 codon 72 polymorphism with HPV was associated with oral carcinoma susceptibility (Kitkumthorn et al., The present study had several limitations. Firstly, only studies written in English or Chinese were included in the meta-analysis. This means that eligible studies published in other languages may have been overlooked, which may have introduced selection bias. Secondly, the sample size of some studies was limited and the results should be interpreted carefully. Thirdly, this study had statistical heterogeneity, although this is extremely common in meta-analyses of genetic association studies. We thus conducted subgroup analyses to identify all factors that contribute to the heterogeneity. Finally, other factors such as the age, gender, life-style that may affect the interaction of TP53 codon 72 polymorphism with oral carcinoma could not be analyzed due to the lack of original data.In summary, this meta-analysis suggests that there is no statistical association between TP53 codon 72 polymorphism and oral cancer susceptibility, independent of tobacco and/or alcohol use and HPV-infection status. However, this conclusion should be confirmed by multi-center and large-scale studies based on multiple ethnic groups.Y-ML, JS, X-TZ, and R-HL conceived and designed the experiments. Y-ML performed the experiments. X-HY, CH, and X-WJ analyzed the data. X-WJ, Y-DY, C-JW, E-MZ, and Z-XY contributed reagents, materials, analysis tools. Y-ML wrote the paper. Y-ML, and X-TZ methods analysis.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.


The association of age at menopause with endometrial cancer remains controversial. Therefore, we quantitatively summarized the evidence from observational studies with a meta-analysis.2. Metaregression was used to explore the potential sources of between-study heterogeneity. Egger's test was used to estimate publication bias. We searched PubMed, Web of Science, Embase, Medline, Chinese National Knowledge Infrastructure (CNKI), and Wan Fang Med online up to March 2019, and all eligible case-control and cohort studies were included in the study. Pooled relative risks (RRs) with 95% confidence intervals (CIs) were calculated using the random-effects model. The dose-response relationship was assessed by restricted cubic spline model. The heterogeneity among studies was evaluated by I Eighteen articles including 957242 subjects with 4781 cases were included in the meta-analysis. The pooled RR (95%CI) of endometrial cancer for the highest versus the lowest age at menopause was 1.89 (95%CI: 1.58-2.26). For dose-response analysis, a nonlinear relationship was found between age at menopause and endometrial cancer, and the positive association became statistically significant when age at menopause was greater than 46.5 years old. This meta-analysis suggested that age at menopause was positively associated with endometrial cancer. For women whose menopausal age over 46.5 years old, the risk of endometrial cancer increased with the age at menopause. Endometrial cancer is the most common gynecological tumor of the female . GloballMenopause as the terminus of women reproductive life is generally defined as a stop of menstruation for a consecutive year . The ageIn order to explore the association between age at menopause and the risk of endometrial cancer, a large number of epidemiologic studies have been conducted , 26–42. Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) guidelines were adopted .∗ age” or “the age of postmenopause” or “age at climacteric” or “climacteric age” or “pausimenia age”) and “endometrial cancer” . Relevant references within included studies were also manually searched. The detailed steps of the literature selection were shown in  We used extended computer-based searches to obtain available studies published in English or Chinese from the databases of PubMed, Web of Science, Embase, Medline, Chinese National Knowledge Infrastructure (CNKI), and Wan Fang Med Online. The search terms used to search articles in this study were “age at menopause”  An observational study (cohort or case-control) was published as an original article. (2) The exposure of interest was categorized age at menopause. (3) The outcome of interest was endometrial cancer. (4) There was reported effect size (relative risk (RR) or odds ratio (OR) or hazard ratio (HR) or incidence rate ratio (IRR)) and 95% confidence interval (CI) for the association between age at menopause and endometrial cancer. (5) We selected the most recent study if data from the same population were used in multiple articles.All identified studies were searched and reviewed carefully by two investigators (Yanjun Wu and Wenjun Sun). If the two investigators had different views on the same article, it would be settled by discussing with the third investigator (Dongfeng Zhang).From each eligible article, we extracted the first author's name, country in which the study was performed, publication year, the type of study design, the follow-up duration of cohort study, age range or mean age at baseline, the number of cases, and controls in case-control studies as well as the person-year of cases in cohort studies. We also abstracted the information about age at menopause, RRs  with their 95%CIs for each category of age at menopause, and adjustment factors in each study, menopausal type, the definition of postmenopausal status, and the source of case information.The Newcastle-Ottawa Scale (NOS)  was usedI2 proposed by Higgins and Thompson [The pooled measure was calculated as the inverse variance-weighted mean of the natural logarithm of RR with corresponding 95% CI to assess the strength of association between age at menopause and the risk of endometrial cancer. Heterogeneity among studies was assessed by Thompson . MetaregThompson . The infThompson .p value for nonlinearity was calculated by testing the null hypothesis that the coefficient of the second spline is equal to 0. The details of the statistical method have been described elsewhere [For dose-response analysis, a two-stage random-effects dose-response meta-analysis  was perflsewhere .P values) were two-sided, and P values less than 0.05 were considered statistically significant.All statistical analyses were performed with StataV.15.0 . All reported probabilities  concerning the interests. Ultimately, 18 articles –42 were After using NOS to assess the quality of the 18 articles included in this study, the mean Newcastle-Ottawa score was 7.8 (range from 6 to 9) for case-control studies and 8.1 (range from 7 to 9) for cohort studies. The detailed results of the quality assessment were summarized in Tables In the 18 articles, nine articles were case-control studies , 38, 41 I2 = 45.0%, Pfor  heterogeneity=0.021)  .Pnonlinearity < 0.01). The positive association became significant when age at menopause was greater than 46.5 years old. The RRs (95% CIs) of endometrial cancer risk were 1.04 (1.03-1.06), 1.17 (1.14-1.20), 1.57 (1.45-1.71), and 2.08 (1.80-2.39) for 47, 50, 54, and 57 years old of age at menopause, respectively  were 1.60 (1.37–1.86), 3.06 (1.64–5.72), and 2.19 (1.74–2.75) for North America, Asia, and Europe, respectively. In the subgroup analysis by study design, the pooled RRs (95%CIs) for case-control studies and cohort studies were 1.80 (1.36–2.38) and 1.98 (1.56–2.52), separately. In the subgroup analysis performed according to menopausal type, the pooled RRs (95%CIs) were 1.72 (1.09-2.71), 2.25 (1.31-3.85), and 1.94 (1.63-2.30) for studies among natural menopausal women, studies among both natural menopause and surgical menopause women, and studies without relevant information, respectively. In the subgroup analysis by reference group of menopausal age, the pooled RR (95%CI) was 2.06 (1.51-2.81) for studies that used age at menopause ≤45 (<45 or ≤ 45 or ≤40) as reference group and the pooled RR (95%CI) was 1.79 (1.52-2.10) for studies that used age at menopause ≤50 (<50 or ≤ 50 or ≤47) as the reference group. In the subgroup by the Newcastle-Ottawa score, the pooled RRs (95%CIs) were 1.67 (1.12-2.49) and 1.98 (1.66-2.37) for studies with a score of 6 or 7 and studies with a score of 8 or 9. The detailed results of subgroup analysis were summarized in p = 0.727), study design (p = 0.623), mean age of study participants (p = 0.538), menopausal type (p = 0.621), reference group of menopausal age (p = 0.554), and the Newcastle-Ottawa score (p = 0.362) showed that these covariates had no significant impact on the heterogeneity. But continent where the study was conducted (p = 0.014) was found to have an influence on the between-study heterogeneity.Multivariable metaregression with the covariates of publication year (p=0.373). The results of the funnel plot were shown in No study had excessive influence on the pooled RR for the highest versus the lowest age at menopause in the influence analysis. No evidence of significant small-study effect for the analyses was found by the visual inspection of the funnel plot and Egger's test , 3.06 , and 2.19 , respectively.The issue of between-study heterogeneity should be paid particular attention in meta-analyses , and invThe study has many virtues. First, compared with the original individual study, our meta-analysis has a large number of included participants, which can make the results more precise and more reliable. Second, the quality of the included articles was relatively high with Newcastle-Ottawa score ranging from 6 to 9. Third, the results are still meaningful after subgroup analysis by continent, menopausal type, the reference group of menopausal age, and Newcastle-Ottawa score, and subgroup analysis by study design also showed the positive association between age at menopause and endometrial cancer in cohort studies. Finally, we conducted a dose-response analysis to explore the association between age at menopause and endometrial cancer quantitatively.However, there are also several potential limitations in our study. First, the adjusted confounders are different among studies and the residual confounding could not be eliminated thoroughly. Second, the age range of participant is disparate in every study and years of follow-up in cohort studies are diverse. Finally, the definition of postmenopausal status and the stratification of age varied among studies, which might affect the result.Results from this meta-analysis indicated that age at menopause was positively associated with the risk of endometrial cancer. When the menopausal age of women exceeded 46.5 years, the risk of endometrial cancer increased with her menopausal age. For these women, they should develop good lifestyles to reduce the risk of endometrial cancer. More well-designed prospective studies are needed to confirm the association in the future.


Background/Aims: Previous results on the association between MTR gene A2756G polymorphism and PCa risk are inconclusive.in-silico tools were utilized to assess this variant.Methods: We used odds ratios (ORs) with corresponding 95% confidence intervals (95% CIs) to evaluate the correlation between MTR A2756G polymorphism and risk of PCa in meta-analysis. Serum expression of MTR was detected by ELISA and Results: Our study included 2,921 PCa patients and 3,095 control subjects. The results indicated that the MTR A2756G polymorphism is linked with an increased risk of PCa using three genetic models . Stratified analysis produced similar results. A significant association was also indicated in advanced PCa from the meta-analysis. Finally, our experiments showed evidence that serum MTR levels in PCa patients with AA genotypes were statistically higher than in those with GG/GA genotypes.Conclusions: Our present study suggests that the MTR A2756G polymorphism may contribute to the risk of developing PCa, particularly in Asian and hospital-based studies. Moreover, serum MTR might be utilized in diagnosis of PCa. Prostate cancer (PCa) remains the most common carcinoma in males and the sixth leading cause of tumor-associated deaths in Western countries . AlthougPrevious studies have indicated that folate-metabolizing genes play a significant role in carcinogenesis through DNA methylation –11. AberInitially, 76 articles were retrieved according to the selection strategy . Among tPheterogeneity = 0.078, P = 0.008, I2 = 47.2  and recessive genetic model . In stratified analysis by tumor stage, we indicated a positive correlation between the MTR A2756G variant and advanced PCa , but not in localized PCa  .In the pooled analysis of all selected studies, an increased risk of PCa was associated with the MTR A2756G polymorphism  in three2 = 47.2 . GA vs.  I2 = 0) . FurtherIn-silico analysis showed evidence that MTR expression in PCa tissue was lower than that in normal tissue  and grade  of PCa in our PCa patient study. String online server indicated that MTR gene interacts with numerous genes. The network of gene-gene interaction has been illustrated in P value for HWE in our PCa cases is 0.002. Furthermore, we investigated the serum expression of MTR by ELISA. Serum MTR levels in PCa patients with GG/GA genotypes were statistically lower than those with AA genotypes , 65 (32.5%), 108 (54.0%), respectively. MAF of MTR A2756G in our study participants was 0.298, which is a bit higher than the one reported for AFR population (0.284) and lower than the one reported for EAS population (0.321). P = 0.305; GG vs. AA: t = 0.92, P = 0.399; GA vs. AA: t = 1.24, P = 0.270; GG+GA vs. AA: t = 1.28, P = 0.257; GG vs. GA+AA: t = 0.83, P = 0.445  may cause low expression of MTR, which means that serum concentration of MTR may be utilized as a detector in the diagnosis of PCa. To further ascertain the effect of MTR A2756G variant in PCa susceptibility, an online gene expression mini-database was utilized to evaluate the expression of MTR in PCa tissues. The Cancer Genome Atlas (TCGA) projects produced large-scale RNA sequencing data and may provide a chance for integrated expression analysis for all genes in large quantities of tumor and normal tissue specimens. It showed that MTR expression in PCa tissue was lower than that in normal tissue, which was consistent with the result from our meta-analysis. In addition, interactions of MTR with other gene might play a role in the current genetic polymorphism. As was shown in P value for HWE in our PCa cases is 0.002, which indicated that the departure of frequencies of MTR gene might exist.Although this is the first study that assesses the association between MTR A2756G polymorphism and PCa risk up to now, some limitations of the present analysis should be noted. First, all enrolled articles were the type of retrospective case-control studies, so selection bias may be existed in the present meta-analysis. Second, it is possible that gene-gene interactions influence the function of gene product and predominate in the development of cancer, which may lead to different ORs. We tried to evaluate the potential interactions, however, raw information from enrolled articles was insufficient. Some more precise researches need to be carried out if other covariates including PSA level, smoking exposure, and so on becomes available in the future case-control studies. Third, no enrolled article was carried out in African population, which should be considered in the future studies. Furthermore, age has been indicated to be associated with PCa risk, we tried to further assess the possible linkage between aging and MTR A2756G polymorphism, nevertheless, the raw data was not available either. Our PCa patients study indicated that serum MTR levels in PCa patients with AA genotypes were statistically higher than in those with GG/GA genotypes. Then, results from meta-analysis indicated a significant association between this variant and advanced PCa. Unfortunately, we have not indicated a positive association between the MTR polymorphism and different stage or grade of PCa in our case study. The possible reasons include: (a), the sample size of our experiment was not large enough; (b), there may be selection bias. In addition, MAF of MTR A2756G in our study participants was a bit higher than the one reported for AFR population (0.284) and lower than the one reported for EAS population (0.321). Meta-analysis is considered to be an effective method for combining previously published results to draw more comprehensive conclusions. In spite of this, some other limitations should be addressed. To start with, only 7 studies based on MTR A2756G polymorphism and PCa were included, which means that the total participant count remains relatively low for comprehensive analysis. Second, only articles written in Chinese or English language were screened; hence, research published in other languages may be neglected. Third, some other risk factors, including family history, lifestyle, and environmental factors were not assessed due to insufficient data. Therefore, future research ought to collect and evaluate this information in more detail. Despite these limitations, some advantages in the present study should be acknowledged. First, heterogeneity is of great significance in meta-analysis. No obvious heterogeneity was detected in the current analysis for MTR A2756G polymorphism. Second, the quality of the selected studies completely satisfied the selection criteria and the statistical efficiency was enhanced. Third, most of the included studies, except one, are consistent with HWE . Fourth,In summary, our current study indicated that the MTR A2756G polymorphism was not only associated with increased PCa risk, particularly in people of Asian descent and in HB studies, but also related to PCa prognosis. Furthermore, studies with larger sample size should be continued to fully elucidate this issue.PubMed, Embase, CNKI, and WanFang databases searches were conducted using the following keywords: ' methionine synthase ' or ' MTR ', 'prostate cancer' or 'prostate carcinoma' and 'variant' or 'polymorphism' . References cited in the retrieved articles were also screened by a manual search. The inclusion criteria were: (a) study of association between MTR A2756G polymorphism and PCa susceptibility or prognosis; (b) available data for all genotypes ; (c) case–control study with no duplicated design.The following criteria were used to screen the eligible studies: (a) evaluation of MTR A2756G polymorphism and PCa susceptibility; (b) containing available data for genotype frequencies; (c) sufficient information for evaluating ORs; (d) case-control studies. Major exclusion criteria were: (a) no available genotype frequency; (b) studies used a case-case or case-only type; (c) If authors published two or more articles using the same population, the article with most complete information was enrolled. Two investigators independently screened all the searched data from studies according to the selection criteria. Any discrepancies were fully discussed by all authors unto consensus. The following parameters from the selected studies were extracted: the first author's surname, publication year, country, race, methods of genotyping, source of control , sample size of the participants, Minor Allele Frequency (MAF) in case and control, Average age for case group and control group. Hardy-Weinberg equilibrium (HWE). In addition, the selected studies contained data on clinical stage of PCa, which were separated into two sub-groups: localized (Gleason score =< 7 or tumor stage T1-T2c) and advanced PCa (Gleason score > 7 or tumor stage T3-T4). Moreover, we used the Newcastle-Ottawa Scale (NOS) score, ranging from zero stars (worst) to nine stars (best), to assess the quality of each study. The NOS score was calculated by comparability of participants and ascertainment of exposure to risks . An artiWe collected the blood in standard cubes without anticoagulant, then utilized serum separator tube (SST) and allow samples to clot for two hours at room temperature or overnight at 4°C before centrifugation for 15 minutes at 1000 ×g. Removed serum and assay immediately or aliquot and stored samples at -20°C or -80°C. Repeated freeze-thaw cycles should be avoided. Serum expression of MTR levels were measured by ELISA kit (CUSABIO Co. ltd.). Micro plate reader capable of measuring absorbance at 450 nm, with the correction wavelength set at 540 nm or 570 nm. The absorbance is related to the standard curve.http://gemini.cancer-pku.cn/) [http://string-db.org/).The online gene expression mini database was utilized to evaluate the expression of MTR in prostate tissues (pku.cn/) . This daZ-test was used to determine the statistical significance of ORs. Chi-square-based Q-test was utilized to evaluate the heterogeneity assumption. A random effects model (DerSimonian-Laird method) was used when the P-value for the Q-test was less than 0.05 [I2 was also utilized to measure the heterogeneity. If I2 > 50%, this would suggest statistically significant heterogeneity among the studies. For the MTR A2756G genetic variant, the association with PCa risk was evaluated according to the models as follows: allelic comparison , homozygote comparison (GG vs. AA), heterozygote comparison (GA vs. AA), dominant model (GG+GA vs. AA) and recessive model (GG vs. GA+AA). Begg's [P values less than 0.05 were considered statistically significant. We calculated the departure of frequencies of MTR gene variant from expectation under HWE by utilizing the Pearson chi-square test. Stata software  was selected to investigate all statistical tests.We utilized odds ratios (ORs) with 95% confidence intervals (CIs) to investigate the strength of association between MTR A2756G polymorphism and PCa susceptibility. han 0.05 , which dhan 0.05 . I2 was . Begg's  and Egge. Begg's  tests weGenotyping of MTR A2756G variant was performed utilizing various techniques in different studies, including quantitative real-time PCR, restriction fragment length polymorphism PCR (PCR-RFLP), and allele-specific Taqman probes and primers as mentioned by Ulvik . In our Overall, 200 newly diagnosed PCa patients from February 2013 to July 2017 in the third Affiliated Hospital of Nantong University and Affiliated Changzhou No.2 People's Hospital of Nanjing Medical University were involved in our study . The age


Whether serum magnesium levels were lower in patients with lung cancer than that in healthy controls is controversial. The aim of this study was to identify and synthesize all citations evaluating the relationship between serum magnesium levels and lung cancer.We searched PubMed, WanFang, China National Knowledge Internet (CNKI), and SinoMed databases for relevant studies before December 31, 2017. Two authors independently selected studies, extracted data, and assessed risk of bias.I2 = 99.6%, P < 0.001) found. Negative associations were found among Asian populations  and European populations . No publication bias was found using the test of Egger and funnel plot.Eleven citations comprising 707 cases with lung cancer and 7595 healthy controls were included in our study. Serum magnesium levels were not significantly lower in patients with lung cancer  when compared to health controls, with significant heterogeneity (Our study suggested that serum magnesium levels had no significant association on lung cancer risk. Lung cancer is the leading cause of death from cancer, resulting 1.38 million people deaths each year . PreviouMagnesium is one of the trace elements in our bodies, and to date, some papers had been published to investigate the association between serum magnesium levels and lung cancer risks. Two papers , 12 repoA comprehensive literature search was conducted in platforms of PubMed, WanFang, China National Knowledge Internet (CNKI), and SinoMed databases up to December 31, 2017. Free words adopted were as follows: “magnesium” or “Mg” combined with “lung cancer” or “lung carcinoma” without restrictions. Reference lists of the studies retrieved were also examined to find any additional study potentially unidentified. The course of study selection was completed by two investigators independently. Any resulting discrepancies were resolved by a third reviewer.The inclusion criteria were as follows: (i) having a prospective design or a case-control design or a cross-sectional study; (ii) evaluating the association between serum magnesium levels and risk of lung cancer; (iii) reporting mean and standard deviation (SD) of magnesium levels (or sufficient data to compute them) both in lung cancer patients and healthy controls; and (iv) studies published in English language or Chinese language. If more than one article referred to the same populations, only the study that included the most lung cancer cases or the latest publication was included.Two investigators independently extracted the following data: (1) first author’s last name; (2) publication year; (3) study design; (4) country; (5) number of lung cancer cases and participants; (6) sex of cases; (7) age range or mean age of the cases; (8) mean and SD of magnesium levels both in lung cancer patients and healthy controls; and (9) method used for detection of magnesium. Any resulting discrepancies were resolved by a third reviewer.The methodological quality of studies was evaluated independently by two researchers using the Newcastle-Ottawa Quality Assessment Scale . The thrI2 and Q tests.   for the pooled sensitivity, suggesting high heterogeneity in the sample of studies. Univariate meta-regression was then carried out to determine the reason of heterogeneity. However, there were no significant contributions about publication year, case number, geographic location, sex, and different methods on this high between-study heterogeneity.Meanwhile, P = 0.586).Sensitivity analysis conducted while removing one study at the time revealed that no single study had essential effect on the whole result. Figure In this study, we assessed the association between serum magnesium levels and risk of lung cancer. We did not find a positive association between serum magnesium levels and lung cancer risk. Through our subgroup analysis, we further found no significant association among Asian and European populations. Significant heterogeneity between studies observed in this meta-analysis should be considered as a major limitation of these findings; however, heterogeneity was mainly related to strength of the association rather than the direction of risk estimate, suggesting overall promising findings on the outcome investigated in the present study.Two previous prospective cohort studies concluded that higher category of dietary magnesium intake had no significant association on lung cancer risk among German population and China population , 29. HowThe existence of heterogeneity among the studies, which is common in meta-analyses , may affSome advantages existed in our study. Firstly, a comprehensive literature search was performed to investigate the relationship between serum magnesium levels and lung cancer risk. Secondly, most of the included studies involved large numbers of patients and healthy controls, and this may strengthen the power of the pooled results. Thirdly, there was no significant publication when tested by Egger and funnel plot, which indicates that our results are stable.The present study has some limitations. Firstly, the individual studies may have failed to control for potential confounders, which may introduce bias in an unpredictable direction. Secondly, ten of 11 studies were from Asia, and 9 were from China, and thus, more related researches from other countries are wanted to verify the association between geographic location and lung cancer risk.Based on the obtained results, we concluded that serum magnesium levels may have no significant association in patients with lung cancer. As we experienced some limitations in our study, such as more studies were from Asia, further studies are wanted to confirm this finding.


Objectives: To carry out an integrated and stratified meta-analysis on occupational exposure to electromagnetic fields (EMFs), metals and pesticides and its effects on amyotrophic lateral sclerosis (ALS) and Parkinson’s and Alzheimer’s disease, and investigate the possibility of publication bias. Methods: In the current study, we updated our recently published meta-analyses on occupational exposures in relation to ALS, Alzheimer’s and Parkinson’s disease. Based on 66 original publications of good scientific epidemiological standard, according to the Meta-analysis of Observational Studies in Epidemiology (MOOSE) and the Grading of Recommendations, Assessment, Development and Evaluations (GRADE) guidelines, we analysed subgroups by carrying out stratified meta-analyses on publication year, statistical precision of the relative risk (RR) estimates, inspection of the funnel plots and test of bias. Results: Based on 19 studies the weighted RR for occupational exposure to EMFs was 1.26  1.07–1.50) for ALS, 1.33 (95% CI 1.07–1.64) for Alzheimer’s disease and 1.02 (95% CI 0.83–1.26) for Parkinson’s disease. Thirty-one studies concerned occupational exposure to pesticides and the weighted RR was 1.35 (95% CI 1.02–1.79) for ALS, 1.50 (95% CI 0.98–2.29) for Alzheimer’s disease and 1.66 (95% CI 1.42–1.94) for Parkinson’s disease. Finally, 14 studies concerned occupational exposure to metals and only exposure to lead (five studies) involved an elevated risk for ALS or Parkinson’s disease and the weighted RR was 1.57 (95% CI 1.11–2.20). The weighted RR for all the non-lead exposures was 0.97 (95% CI 0.88–1.06). Conclusions: Exposure to pesticides increased the risk of getting the mentioned neurodegenerative diseases by at least 50%. Exposure to lead was only studied for ALS and Parkinson’s disease and involved 50% increased risk. Occupational exposure to EMFs seemed to involve some 10% increase in risk for ALS and Alzheimer’s disease only. Parkinson’s disease, amyotrophic lateral sclerosis (ALS) and Alzheimer’s disease are the most common neurodegenerative disorders. They primarily affect ageing individuals and are characterized by a steady progressive course because of increasing loss of specific neurons in the brain. Over the last two decades our understanding of the pathophysiological processes of neurodegeneration has grown extensively owing to huge progress in biochemical research. Targets in all three diseases are specific large molecules that are essential for ensuring normal function and survival of the neuron cell. There is a continuous turnover of these molecules and several enzymatic pathways are involved in their production and their degradation. With age, more and more damaged or misfolded defective molecules are stored in inclusion bodies (“garbage bags”) in and between neurons, thus enhancing degeneration of cells. The defective molecules also disturb the function of the neuron, leading to cell death. Promoters in the molecule-damaging process are hereditary insufficiencies as well as exposure to chemicals, all influencing the turnover of the target protein. However, the damaging process can also be influenced by stress and degree of metabolic activity in the cell, as well as other factors.Several systematic literature reviews including meta-analyses have been published on epidemiological studies of occupational exposure and neurodegenerative diseases. The most frequently studied exposures have been pesticides ,4,5,6,7,The aim of the present paper was to update our database, carry out integrated and stratified meta-analyses, based only on publications of good scientific standard, on neurodegenerative diseases in relation to main occupational exposures, and investigate the possibility of publication bias.Using the same search criteria as in our previous meta-analyses ,20,21 anRisk estimates from the selected studies are reported as relative risks (RRs), as the outcome is rare. For the purposes of this research, we considered odds ratio (OR) and hazard ratio (HR) equivalent to RR. Where a study reported unadjusted as well as multivariable-adjusted risk estimates, we only considered the adjusted estimates. Studies that reported stratified estimates for sex were treated as separate studies and the estimates of each were included. Where exposure was categorized into different levels the risk rate for the highest level was used. Some studies reported estimates for more than one neurodegenerative disorder and all are included in the summary.I2 statistic and applied the recommended cut-offs of 25%, 50% and 75% degrees of heterogeneity. We stratified the data by study characteristics , selected a priori, and used meta-regression to evaluate the significance of the stratification variable. Based on the I2 criterion and the meta-regression, a random effects model was found to be the most appropriate choice in all of our analyses; hence, the results are reported only with random effects estimates. The weights used for pooling the risk estimates were equal to the inverse-variance weighting. Pooled risk estimates are presented with 95% confidence intervals (CIs).We examined the fixed effects model as well as the random effects model by considering statistical heterogeneity. To this end, we used the elog(RR), the essential factor that affects the width of the 95% CI for RR. The practical application was to use a value of 0.26 for this standard error, which rendered the two strata (low vs. high precision) to be approximately equal in numbers of estimated RRs for several of the comparisons. The width of the CI is dependent both on the total number of subjects in the study and on the number of cases with the specified diagnosis. We therefore use the phrases “estimate with high precision” and “estimate with low precision” rather than “study with high precision” and “study with low precision”. It is possible that the same study with more than one diagnosis analysed may have contributed to one estimate with high and another estimate with low precision.Publication bias was analysed by inspection of the funnel plot, in which the estimates of RR should be distributed symmetrically around the weighted RR unless the publication was affected by bias. The rank correlation test proposed by Begg and Mazumdar  was usedwww.Stata.com, StataCorp, College Station, TX, USA), and described in articles from the STATA journal [Statistical analyses were conducted using procedures for different aspects of meta-analyses available in STATA software, version 15.2  and contributed data to meta-analyses of risk rates regarding exposure to EMFs: eleven studies on ALS ,35,36,37To test the hypothesis that publication bias had a higher influence on earlier publications we made a Metacum on ALS and Alzheimer’s disease together, stratified by year of publication See . The weiThe corresponding funnel plots of studies on ALS and Alzheimer’s disease together see  showed aEstimates from the studies on ALS and Alzheimer’s disease were grouped into estimates with narrow CIs (high precision estimates) and estimates with wider CIs (low precision estimates) and a Metacum based on an ordering by precision is shown in Thirty-one studies fulfilling good scientific standards (Armon class II–III) contributed data to meta-analyses of risk rates for pesticide exposure: five studies on ALS ,45,46,47p = 0.060).The weighted RR for ALS, Alzheimer’s and Parkinson’s disease was 1.35 (95% CI 1.02–1.79), 1.50 (95% CI 0.98–2.29) and 1.66 (95% CI 1.42–1.94), respectively. The weighted funnel plot see  showed ap = 0.026. The weighted RR for the six Parkinson’s studies published before 2005 was 1.98 (95% CI 1.41–2.77), compared with 1.57 (95% CI 1.33–1.85) for the 18 studies published in or after 2005. The estimates from these studies were also dichotomized according to precision. The twelve estimates with the highest precision, in studies [Only analysing the 24 out of 31 studies concerning Parkinson’s disease resulted in even more marked publication bias, Begg’s test 95% CI 1.–1.85 for studies ,68,71,72 studies ,67,69,70p = 0.70.Fourteen studies fulfilling good scientific standards (Armon class II–III) contributed data to meta-analyses of risk rates regarding exposure to metals: six studies on ALS ,74,75,76Exposure to welding involves exposure to both a mixture of metals and EMFs. Six studies of high scientific standard reported risk estimates for ALS ,78,82,83Recently we have presented three meta-analyses on occupational exposures in relation to ALS, and Alzheimer’s or Parkinson’s disease ,20,21, bElevated weighted RRs were obtained for only ALS and Alzheimer’s disease. In almost all the included studies, occupational exposure to low-frequency EMFs was assessed from a validated job exposure matrix applied to job titles according to census registers or other registers. Questionnaires and/or interviews were used in two studies only ,40. StudThe weighted RR from studies published in 2005 or later was 1.12 (95% CI 1.04–1.20) and there was no publication bias for this period. The weighted risk estimate for earlier publications (RR = 1.53) was hampered by an evident publication bias indicating that the true risk-RR was closer to 1.12. This risk estimate also corresponds to the estimate derived from the high-precision studies where the weighted RR was 1.14.Below we discuss four systematic reviews and meta-analyses on exposure to EMFs and neurodegenerative diseases. Only some of the included studies fulfil good scientific standards, according to Armon   or a work place cohort with specified exposures. Furthermore, the dosage of exposure is not treated uniformly. In some studies exposure is graded Yes or No, in others the dosage is graded into 3–4 levels (but not uniformly between studies). Based on this and other heterogeneities in exposure we selected risk estimates covering medium or high levels and also gave priority to many years of exposure instead of peak exposures. The risk estimates given should thus be interpreted as averages for recurrent occupational exposures. However, according to the decision protocol (see Appendix) we meticulously have scrutinized the quality of exposure data in every included publication.Our meta-analyses showed that exposure to pesticides carried at least a 50% increased risk for contracting Parkinson’s disease, Alzheimer’s disease or ALS, ranked in order of risk. Exposure to lead seemed to involve at least 50% increased risk for getting ALS or Parkinson’s disease, while non-lead exposures did not seem to involve risk. Occupational exposure to EMFs seemed to involve some 10% increase in risk for ALS and Alzheimer’s disease, but no such indication of risk was found for Parkinson’s disease.


Background: Meta-analysis is a widely used tool in which weighted information from multiple similar studies is aggregated to increase statistical power. However, the exponential growth of publications in key areas of medical science has rendered manual identification of relevant studies increasingly time-consuming. The aim of this work was to develop a machine learning technique capable of robust automatic study selection for meta-analysis. We have validated this approach with an up-to-date meta-analysis to investigate the association between diabetes mellitus (DM) and new-onset atrial fibrillation (AF).Methods: The PubMed online database was searched from 1960 to September 2017 where 4,177 publications that mentioned both DM and AF were identified. Relevant studies were selected as follows. First, publications were clustered based on common text features using an unsupervised K-means algorithm. Clusters that best matched the selected set of potentially relevant studies (a “training” set of 139 articles) were then identified by using maximum entropy classification. The 139 articles selected automatically on this basis were screened manually to identify potentially relevant studies. To determine the validity of the automated process, a parallel set of studies was also assembled by manually screening all initially searched publications. Finally, detailed manual selection was performed on the full texts of the studies in both sets using standard criteria. Quality assessment, meta-regression random-effects models, sensitivity analysis and publication bias assessment were then conducted.Results: Machine learning-assisted screening identified the same 29 studies for meta-analysis as those identified by using manual screening alone. Machine learning enabled more robust and efficient study selection, reducing the number of studies needed for manual screening from 4,177 to 556 articles. A pooled analysis using the most conservative estimates indicated that patients with DM had ~49% greater risk of developing AF compared with individuals without DM. After adjusting for three additional risk factors i.e., hypertension, obesity and heart disease, the relative risk was 23%. Using multivariate adjusted models, the risk for developing AF in patients with DM was similar for all DM subtypes. Women with DM were 24% more likely to develop AF than men with DM. The risk for new-onset AF in patients with DM has also increased over the years.Conclusions: We have developed a novel machine learning method to identify publications suitable for inclusion in meta-analysis.This approach has the capacity to provide for a more efficient and more objective study selection process for future such studies. We have used it to demonstrate that DM is a strong, independent risk factor for AF, particularly for women. Meta-analysis is a powerful epidemiological tool that is increasingly used in all fields of scientific research. It seeks to amplify statistical power by aggregating weighted information from multiple similar studies  is the most commonly sustained arrhythmia that is associated with substantial morbidity and mortality  guidelines  where an association between DM and AF was reported. Patients with established comorbidities such as cardiac disease, hypertension and obesity/body mass index (BMI) were included so that the potential influence of these conditions on AF could be evaluated and the adjusted risk ratios were calculated. To reduce publication bias, previous review and meta-analysis papers were excluded. Furthermore, where multiple findings were reported from the same or overlapping patient datasets, the most contemporary study was used.R, an open-source environment for data mining, text processing, machine learning and statistical analysis  was used to fit a predictive model to the labeled training set and this was applied across all clusters to identify clusters of articles that best matched this selection Figure . Titles To validate our machine learning assisted screening approach, study selection using this approach was compared to the results of conventional manual selection , year of subject's enrolment, country of the study, number of subjects, AF subtypes, DM subtypes, sex, age, year of follow up and comorbidities. Quantitative estimates of the association between DM and AF were also extracted from the original publications together with their respective confidence intervals (CIs). These included either hazard ratios (HR) for cohort/randomized studies or odds ratio (OR) for retrospective case-control studies. Age-and/or-gender/none adjusted  estimates were extracted from individual studies as well as other multivariate adjusted estimates when provided. Relative risk (RR) was estimated using the DerSimonian and Laird random-effects model across different studies and subgroups such as men vs. women.The most conservative estimates provided in individual studies were used in the meta-analysis to take advantage of all included studies. Multivariate adjusted estimates were applied wherever possible using a subset of these studies by utilizing the studies with age, sex, and additional multiple adjustments for various reported risk factors. Furthermore, risk estimates for AF were adjusted separately for hypertension, obesity and various heart diseases in addition to other adjustments.T-test. Publication bias was assessed using the Egger regression test and is presented as a funnel plot.The effects on RR of CIs, publication year, year of the study, age, mean follow-up years and number of adjusted factors were investigated using polynomial regression analysis with Pearson's correlation. Throughout this study, statistical significance was assessed using the 2-tailed The initial PubMed search yielded a total of 4,177 publications and these were automatically sorted into 14 groups using unsupervised clustering. The training dataset consisted of 139 articles of which 26 were labeled as potentially relevant and 113 not relevant Figure . Manual /modified Jadad score) ranged from 5 to 9 for the cohort/randomized/case-control studies . The overall average score was 7.4 . Funnel plot assessment .Analysis of the combined cohort, randomized and case-control studies identified that patients with DM had ~49% greater risk of developing AF  compared to individuals without DM Figure . The fort Figure  providedNine studies were adjusted for risk estimates based only on age-and/or-gender/none, while 20 included adjustments for multiple risk factors including different combinations of hypertension, BMI, height, smoking, blood pressure, alcohol consumption, various cardiac diseases and race Figure . The levI2 statistic) included adjustments for all the three factors and the estimated overall risk of AF in patients with DM was lower   Figure .p = 0.8), and mean follow-up duration displayed a minor impact .The effects on RR of the width of the CIs (number of enrolled patients), publication year, study location, year of patient enrollment, AF subtype, DM subtype, sex, age and mean follow-up year were also investigated. Continental location Figure  had no st Figure . Furthert Figure  (narrowep = 0.5) between the risks of the different AF subtypes  in patients with DM, compared with all subtypes of AF . Similarly, using the multivariate risk model, no significant difference in the AF risk estimate (p = 0.4) was observed in the studies grouped by undefined DM subtypes , DM type 2 , and DM type 1  . Our results with multivariate adjustment for confounders found no significant difference p = 0. between p < 0.001) .Our results with multivariate adjustment for confounders show a higher risk of AF in women  compared to men  Figure . The RR This systematic review and meta-analysis systematically analyzed 29 studies with a total of 8,037,756 participants from three continents, selected from 4,177 articles returned from an initial strategic search. To our knowledge, this is the largest study of this kind to explore the association between AF and DM, and the first to employ a machine learning approach to facilitate study selection.From our literature research, the total number of publications related to the association investigated here has increased more than 4-fold over the last 10 years. The growth in medical research publications is accelerating across the board and we expect that it will no longer be feasible in the near future to maintain current manual study selection methods for large-scale meta-analyses and systematic review. Therefore, there is an urgent need to develop an intelligent automated approach, such as machine learning, to facilitate identification and selection of relevant articles for this purpose.In this study, we have used machine learning to assist the screening of potentially relevant articles for large-scale meta-analyses and systematic review. To the best of our knowledge, we are the first to employ such an approach to facilitate study selection. With our approach, the burden of manual screening is reduced from all the articles returned by the initial online strategic search to those in the training set and in the principal cluster(s) identified by supervised machine learning. In this study, the number of publications for which manual screening was needed reduced from 4,177 to 555 using machine learning assisted screening, i.e., an 87% reduction. The iterative clustering approach (unsupervised machine learning) utilized vastly improved the similarity of the articles in individual clusters. Subsequent use of supervised machine learning (maximum entropy classification) with a representative training data subset enabled us to screen articles more rapidly than conventional manual procedures.Automated methods of meta-analyses have been proposed in previous studies; however, few have automated the most labour intensive initial study screening stage. A study by Michelson  proposedA further benefit of machine learning-assisted selection of studies for meta-analysis is that identical screening can be applied and consistent results will be obtained if the meta-analysis is recapitulated with additional articles, minimizing the introduction of time-related bias. Machine learning-assisted screening therefore has the capacity to provide more efficient and more objective study selection for future meta-analyses. We expect that the novel automated approaches demonstrated will be extended and refined in the future.AF is associated with several well-established risk factors including age, hypertension, BMI and various cardiovascular diseases. The RRs assessed for these established risk factors have varied between studies and over time. Recent data from the Framingham Heart Study indicate that mean systolic blood pressures and evidence of left ventricular hypertrophy have declined, likely as a consequence of improved therapy for hypertension. In contrast, increasing BMI and DM have contributed to increased risk of AF in the population  for AF incidence using the most conservative adjustments provided in each individual study. After adjusting for three common risk factors  in addition to age, gender and possible others, the estimated risk for AF is 1.23 (95% CI 1.03–1.46). Our results are consistent with others, e.g., Dublin et al. estimated the RR for AF is 1.40 (95% CI 1.15–1.71) after adjusting for confounders including hypertension and BMI  (Alves-Cabratosa et al., AF subtype was not analyzed in most of the studies. Only one study reported an association between cumulative exposure to DM and new-onset AF with all three AF subtypes (Dublin et al., Patients with type 2 DM were exclusively enrolled in 7 out of the 29 selected studies, and there was only one study on the risk estimate of AF in patients with type 1 DM. Twenty-one studies did not explicitly define DM subtypes in their studies. Our study found no significant risk difference among these sub-groups. Interestingly enough, the study on type 1 DM by Dahlqvist et al.  reportedAF incidence and prevalence are lower in women than in men. Sex hormones and delayed onset of cardiovascular disease in women are likely to contribute to these differences. However, the absolute number of women with AF exceeds that of men because women live longer (Perez et al., The prevalence of AF in the general population is projected to more than double in the next few decades, becoming a global epidemic (Krijthe et al., Nevertheless, there are some important findings that are yet to be confirmed, such as the impact of median follow-up and age from new-onset DM on AF risk estimate. Some individual studies have suggested that the first 4–5 years after new-onset DM is the most vulnerable period for developing AF (Aksnes et al., The recent review paper by Lau and his colleagues has proposed a promising integrated care model that incorporates risk factor management, including DM, as the fourth pillar of AF care alone with rate control, rhythm control and anticoagulation therapy (Lau et al., Our study has several limitations. The utility of a promising machine learning approach for publication selection is demonstrated in this study for the first time. However, future development and validation is needed for this approach to achieve its full capacity. The studies included in this meta-analysis are heterogeneous and include differences in patient demographics and marked variation in follow-up duration, though it is substantially reduced in subgroup studies. Furthermore, not all studies were adjusted for the multiple risk factors known to influence AF incidence. In addition, the efficacy of glycemic control on AF risk has not been assessed. Finally, the AF risk estimate for patients with DM in this study only demonstrates the possible causal association between DM and AF, the exact underlying mechanism remains elusive due to the complex nature of concurrent diseases and the limitations of current population-based clinical studies.We have demonstrated the utility of machine learning for meta-analyses. Our study has indicated that the AF risk estimates in patients with DM may be underestimated and has reinforced the view that DM is an independent risk factor for AF even after adjusting for other known concurrent risk factors.ZX and JZ conceived and designed the experiments; ZX, TL, GT, MG, MS, AG, and JZ data curation; PG, BS, MS, AG, and JZ investigation; ZX, TL, GT, and MG data analysis; All authors draft, reviewing and editing of 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: Several studies including some genome-wide association studies (GWAS) had shown that BAK1 gene rs210138 polymorphisms might be associated with testicular germ cell tumors (TGCT). Here we tried to sum up the association through a systematic review and meta-analysis.Methods: Studies associated with BAK1 rs210138 and TGCT was systematically searched in databases. The effect size was pooled according to ORs and 95% CIs.Results: Our systematic review and meta-analysis comprised 14 articles. Significantly increased risk of TGCT was found in eligible GWAS and follow-up studies, in overall group and its Caucasian subgroup.Conclusions: Compared with adenine (A), BAK1 rs210138 guanine (G) is associated with increased risk of TGCT. Well-planned studies with larger sample size and more subgroups are needed to verify the risk identified in our systematic review and meta-analysis. Testicular germ cell tumor (TGCT) manly consist of seminomas and non-seminomas . The proe testis . Severale testis  had showclinicaltrials.gov, Cochrane Library, PubMed, Embase. Term “rs210138” was used in GWAS Catalog. And in other databases, these terms were used without limitation: “BAK1” AND “cancer of testis OR carcinoma of testis OR testicular cancer OR testis cancer OR ball cancer OR testicular germ cell tumors OR TGCT” AND “polymorphisms OR polymorphism.” The last search update was on Nov 30, 2018. For additional studies, we made a manual search in reviews and references of related studies.Independently, two researchers systematically searched these databases: GWAS Catalog, Wanfang, CNKI, Independently, two researchers made the selection according to the following inclusion criteria: (1) evaluation of the association between BAK1 rs210138 and TGCT susceptibility; (2) case-control study; (3) studies focusing on tissues of human beings; 4) elaborate genotype data in non-GWAS study or enough data in GWAS study could be acquired. Exclusion criteria: (1) duplication of previous publications ; (2) review, comment, and editorial; (3) study without enough data; (4) studies focusing on cell lines. Dissertation thesis were included in the analysis. We took experience and inspiration from the article we have published recently  , two invP < 0.05 as significant except for specified conditions, where a certain P-value was declared.This meta-analysis complied with the PRISMA checklists . Pooled clinicaltrials.gov = 0, CNKI = 22, Wanfang = 3, GWAS Catalog = 5, other sources = 5). In total, we captured 66 articles from databases  , and I–VAbove all, we found BAK1 rs210138 guanine (G) was associated with increased risk of TGCT in most genetic models in the meta-analysis of single case-control studies. In GWAS studies, follow-up studies and their meta-analysis based on OR and 95% CIs, BAK1 rs210138 guanine (G) also showed association with increased risk of TGCT in allelic comparison, which was consistent with the results in the meta-analysis of single case-control studies.Meanwhile, our meta-analysis had several limitations which should be mentioned. Up to now, number of eligible studies for our meta-analysis were small. There was inadequate data for subgroup analyses. Omission of studies in other languages or unpublished studies might happened. In our eligible GWAS and follow-up studies, we could not retrieve elaborate genotype data. With those limitations, the study provided some insights on the potential association between BAK1 rs210138 and TGCT susceptibility.Our results suggested that: Compared with adenine (A), BAK1 rs210138 guanine (G) is associated with increased risk of TGCT. Well-planned studies with larger sample size and more subgroups are needed to verify the risk identified in our systematic review and meta-analysis.The Ethics Committee of the First Affiliated Hospital of Xiamen University approved the study protocol. Written informed consent was obtained from all patients enrolled in the investigation.JQ and JX designed the study, drafted and substantively revised the manuscript, and proof the English language. JQ, YY, and XZ accumulated the data. JQ and YY were for analysis and interpretation of the data. All authors read and approved the final 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.


IL‐10) gene polymorphisms (−1082G/A) in the progression of cardiovascular disease (CVD). Therefore, this study processed a systemic meta‐analysis to verify this association.Previous studies have generated controversial results about the association of interleukin 10 (IL‐10 (−1082G/A) polymorphism and CVDs risk were obtained by searching PubMed and Embase databases. We analyzed the genotype data for meta‐analysis. The results were evaluated by odds ratios (ORs) and 95% confidence intervals (CIs). Meanwhile, our meta‐analysis was also performed sensitivity analyses, heterogeneity test, and identification of publication bias.The publication studies on the IL‐10 (−1082) could increase the risk of CVDs in the 31 case–control studies for all genetic models. .The present meta‐analysis suggested that the risk with allele G is lower than with allele A for CVD. The G allele of IL‐10 (−1082G/A) polymorphism and CVDs risk was significant. This meta‐analysis proposes that the IL‐10 (−1082G/A) polymorphism may serve as a risk factor for CVDs.In genetic models, the association between the  The last report was updated on April 17, 2017. When included in the analysis, qualified researches must meet the following criteria: (1) the association studies about the IL‐10 polymorphism and susceptibility to related CVDs must follow case–control study strategies; (2) all patients meet the diagnostic criteria for CVDs in the candidate studies; (3) There is enough available data to calculate the odds ratio (OR) and 95% confidence interval (CI). The major exclusion criteria for studies were: (a) not a case–control study; (b) studies that have been republished; and (c) no feasibility data studies. This meta‐analysis was performed conforming to the Preferred Reporting Items for Systematic Reviews and Meta‐analyses (PRISMA) guidelines. Finally, data for meta‐analysis were available from 31 studies, including 10,502 cases and 7,865 controls 61  in controls. Different ethnicity was categorized as Asians, Caucasians, and Mixed.Primary investigators used standardization requirements to extract qualified research from the database. They organized the data into tables and submitted them to coauthors for review to ensure that data are strict and reasonable. We collected the following information from the retrieved literature: first author's name, publication year, control source, country and ethnicity of population, the methods of genotype, the number of cases and controls, the distribution of genotype in cases and controls, and the 2.3IL‐10 (−1082 G/A) polymorphisms and CVDs risk was assessed by OR and the corresponding 95% CI for each study. Different ORs (95% CI) were calculated using the following models: the allele model (G vs. A), the dominant model (GG + AG vs. AA), the recessive model (GG vs. AG + AA), the homozygote model (GG vs. AA), and the heterozygote model (GA vs. AA). Using the Cochran’s Q statistic and I2 test to evaluated the statistical heterogeneity between eligible studies, which was considered as significant when pQ＜0.1 or I2 > 50%. A fixed‐effect model (the Mantel–Haenszel method) was adopted if p > .1 and I2 < 50%, and a random‐effect model (DerSimonian–Laird method) was used if p < .1 and I2 > 50%. Furthermore, we conducted subgroup analysis based on ethnicity  and disease subtype  to explore the sources of heterogeneity. All statistical analyses were performed using software STATA 15.0 (StataCorp LP).The 33.1As shown in Figure IL‐10 (−1082 G/A) genotype distributions in the controls from all studies has carried on HWE (Table The characteristics of the studies are listed in Table L‐10 −108 G/A geno3.2IL‐10 (−1082 G/A) polymorphism and CVDs risk, as well as homozygote test and heterogeneity test. The combined results showed that the variant genotypes were related to increase the CVDs risk in different genetic models . The forest plots for −1082 G/A in the allele model are shown in Figure As shown in Tables L‐10 −108 G/A poly3.3p = .000, I2 > 50%; dominant model GG/AG vs. AA: p = .000, I2 ＞ 50%; recessive model GG vs. AG/ AA: p = .000, I2 > 50%; homozygote comparison model GG vs. AA: p = .000, I2 > 50%; heterozygote comparison model GA vs. AA: p = 0.000, I2 > 50%)  Tables  and 4. T3.4p values greater than 0.05 , which providing statistical evidence to the funnel plots’ symmetry.To evaluate the stability of the pooled results, sensitivity analysis was performed. The result showed that there were no substantial changes in ORs after canceling each study Figure . As to t4IL‐10 is a multifunctional cytokine that has been considered as a candidate gene for kinds of CVD  SNP with CVDs risk.Cardiovascular disease is broadly testified as atherosclerosis, underlying vascular aberrations, as well as formation of coronary plaques  in CVD risk. The results showed that there exists a significant relationship between IL‐10 (−1082 G/A) and CVDs risk. This study was conducted by critically evaluating 31 individual case–control studies involving the IL‐10 (−1082 G/A) polymorphism and CVDs risk. We found that the IL‐10 (−1082 G/A) was associated with an increased risk of CVDs among Asians. As far as we know, this updated meta‐analysis includes the largest samples and the most cogent conclusions.For this meta‐analysis, we aimed to identify the dispute about the role of IL‐10 (−1082A/G) were associated with CVDs risk, which was consistent with a present study, as reported by Yang et al.  were not associated with ischemic stroke risk  and ischemic stroke has been found in the Italian population  polymorphism was significantly associated with increasing the cerebral infarction risk in Asians  with CVDs, and the results remain controversial. Although previous individual studies have reported an association, the overall result of the present analysis support the relationship between IL‐10 −1082G/A polymorphism and CVD risk in all genetic models. There are three potential reasons can explain the difference among these findings. First, the relatively small sample sizes of included studies might lead to false positive or negative results. Second, the inconsistent results could result from different genotype frequencies among enrolled subjects, especially in different ethnic groups. Third, due to the complicated pathogenesis of CVDs, it is difficult to explain that the SNP in a single gene could increase risk of CVD without a contribution of other polymorphic susceptibility genes.In our study, we observed that the allele and genotype frequencies of IL‐10 production is regulated by the complicated interaction between gene and environment, the effect of single genetic mutation is limited. So the evidence offered by this meta‐analysis should be accepted with caution. However, our study also exist some advantages: the reasonable‐designed search and selection method could increase the statistical power and the accuracy of the results, at the same time, the results did not show any evidence of publication bias.Limitations also exist in our study. First, the study lack of detailed information in patients. Second, the studies with small sample size (<100 cases and controls) may overestimate the relationship. Third, the origins of heterogeneity may contain many factors, such as the diverse characteristics of the control group and the different methods of diverse genotyping. Finally, the human IL‐10 −1082A/G polymorphism is associated with an increased risk of CVDs. In subsequent studies, we will conduct functional studies to confirm our conclusions.In conclusion, the results of our study indicate that the The authors have no conflict of interest to report.