Many studies have explored the relationship between coffee—one of the most commonly consumed beverages today—and obesity. Despite inconsistent results, the relationship has not been systematically summarized. Thus, we conducted a meta-analysis by compiling data from 12 epidemiologic studies identified from PubMed and Embase through February 2019. The included studies assessed obesity by body mass index  or waist circumference ; analyzed the measure as a continuous outcome or binary outcome. Using random effects model, weighted mean difference (WMD) and 95% confidence interval (CI) were obtained for continuous outcomes; summary relative risk (RR) and 95% CI for the highest vs. lowest categories of coffee intake were estimated for binary outcome. For BMI, WMD was −0.08 ; RR was 1.49 . For WC, WMD was −0.27  and RR was 1.07 . In subgroup analysis by sex, evidence for an inverse association was more evident in men, specifically for continuous outcome, with WMD −0.05  for BMI and −0.21  for WC. Our meta-analysis suggests that higher coffee intake might be modestly associated with reduced adiposity, particularly in men.  Having gained significant popularity worldwide in the last few decades, coffee has been a preferred source of caffeine for many people. Coffee is high in not only caffeine but also other bioactive compounds, such as polyphenol and chlorogenic acid that have been suggested to confer diverse health benefits . Recent Considering that obesity has been established as a major underlying cause for the aforementioned health risks, it is plausible that coffee intake may be associated with reduced risk of obesity. Several studies have examined the effect of coffee intake on adiposity, as assessed by diverse anthropometric measures including body mass index (BMI) or waist circumference (WC) ,13,14,15The design, analysis, and reporting of this meta-analysis were based on the Meta-analysis Of Observational Studies in Epidemiology (MOOSE) checklist . Three aPubMed and Embase were searched for epidemiologic studies published up to February 2019 using the provided search terms . The lanTo be included under the scope of this paper, studies had to be an observational study  or experimental study  that examined the relationship between coffee intake and adiposity as indicated by weight, BMI, or WC. Studies that tested the effect of coffee extract supplements were excluded. We also excluded studies that reported results only from univariable analyses because estimates from these studies are highly likely to be confounded by factors such as age, alcohol intake, and smoking. Because obesity patterns in men and women generally differ, with men being more prone to abdominal adiposity (“apple shape”) and women more prone to lower-body adiposity (“pear shape”), coffee intake, if plays a role in inducing obesity, may exert differential effects by sex. For this reason, we primarily extracted separate estimates by sex unless studies provided only a pooled estimate. The flow diagram of the study selection is presented in From each study, the following information was extracted: First author, publication year, study design, characteristics of study population , unit and type of coffee intake, measure and definition of obesity, and variables controlled for. For studies that used continuous outcome , mean or mean difference and corresponding standard deviation were extracted. For studies that used binary outcome , relative risk (RR)  and 95% confidence interval (CI) were extracted.To obtain summary estimates accounting for potential clinical and methodological variations across studies, we pooled the results using DerSimonian–Laird random effects model . For con2 statistic, which represents the percentage of total variation across studies that is attributable to between-study heterogeneity [2 values of <25%, 25–49%, 50–74%, and ≥75% were considered to represent low, moderate, high, and very high heterogeneity, respectively. Potential for small study effects, such as publication bias, was assessed using Egger’s test [Egger < 0.1 instead of 0.05.Heterogeneity in the relationship between coffee intake and adiposity across studies was quantified by Iogeneity . I2 valur’s test . Given a9 were conducted by a priori selected variables related to potential effect modifiers .Subgroup analyses and meta-regressionFor statistical significance, two-sided α was set at 0.05. All statistical analyses were conducted using STATA/SE V15.0 .A total of 12 studies were included in this meta-analysis ,14,15. D2 = 65%), it was largely driven by difference in effect size and not by inconsistency in the direction of associations. There was no evidence of small study effects, such as publication bias (PEgger = 0.39). By sex, a significant inverse association was found in men  but not in women  (heterogeneity = 0.68).Based on three studies ,8,11 tha2 = 84%) . Heterog2 = 94%) (2 = 86%). Small study effects, such as publication bias, were not indicated (PEgger = 0.19). In subgroup analysis by sex, the association was not significantly heterogeneous (Pheterogeneity = 0.13), but a significant positive association was pronounced in women  than in men   B, with selevant) .2 = 45%) (Egger = 0.98). Within each sex, an inverse association was suggested, but it was statistically significant only in men  (heterogeneity = 0.58).Based on four studies ,10,11,152 = 45%) A. There I2 = 0%) . Heterog2 = 82%), with studies published in Asia showing opposite associations [Egger = 0.92). In subgroup analysis by sex, higher coffee intake was associated with reduced central obesity in men  but increased central obesity in women  (heterogeneity = 0.59).A total of six studies were included in this meta-analysis ,12,13,14ciations ,12,13,145, 1.86) . HoweverIn this meta-analysis of observational epidemiologic studies, higher coffee intake was significantly associated with modestly lower BMI and WC in men but not in women. However, these findings were not replicated in the meta-analysis of observational epidemiologic studies that defined overall or central obesity using specific cutoffs of BMI and WC. For overweight or obesity as defined by BMI, there was no evidence of an association in men, while a higher coffee intake was significantly associated with an increased risk in women. For central obesity as defined by WC, higher coffee intake was associated with reduced risk, albeit statistically nonsignificant, in men; a statistically nonsignificant but positive association was indicated in women.For BMI and WC, respectively, the inconsistent findings between the continuous and binary outcomes can be explained from several perspectives. First, because no study examined coffee intake in relation to both continuous and binary anthropometric measures, different studies contributed to our meta-analyses of continuous and binary forms of a given adiposity measure. Studies differ in diverse factors including study design, population, and methodological quality, and these variations may have contributed to the inconsistency. For instance, studies conducted in Korea where instant coffee mixes containing sugar and creamer account for a major proportion of overall coffee consumption were included in the meta-analyses of binary outcome only, mostly contributing highly positive associations between coffee intake and obesity ,13. AlteDespite the modest degree of association between coffee intake and continuous BMI and WC, there are several biological mechanisms that add support to the anti-obesity benefit of coffee. Biologically active compounds in coffee, such as chlorogenic acid, caffeine, trigonelline, and magnesium, have shown to be associated with anti-obesity benefits . SupplemEgger < 0.1). Although Egger’s test can be low-powered given a small number of studies, Egger’s test is still more sensitive in detecting many types of biases associated with small studies compared with other equivalent tests such as Begg’s test.There are several limitations in our meta-analysis. First, our study extends the limitations of the included studies, which were mostly cross-sectional. Thus, the validity of our meta-analysis to infer temporal relationship between coffee intake and adiposity is compromised; residual confounding by smoking cannot be ruled out. Larson et al. performed a sensitivity analysis among never smokers and an association became attenuated and statistically nonsignificant . AdditioYet, our meta-analyses have strengths as well. To our knowledge, this is the first meta-analysis that summarized the relationship between coffee intake and adiposity. By analyzing coffee intake in relation to both BMI and WC, and in both continuous and binary forms, this meta-analysis exhaustively examined any potential effect that coffee intake may have on both the amount and the distribution of adiposity. As a meta-analysis combines multiple studies, the increased statistical power enabled us to detect a modest benefit of coffee intake in reducing adiposity, as indicated by a small reduction in BMI and WC.In conclusion, coffee intake may be modestly associated with lower adiposity as indicated by BMI and WC, particularly in men. While the benefit of coffee intake against obesity was not as fundamental as balanced diets and physical activity, in view of considerable evidence supporting the protective effect of coffee against major chronic diseases ,3, coffe


To assess the relationship between 5α-reductase inhibitors (5ARIs) and the risk of male breast cancer (MBC).We systematically searched Medline via PubMed, Embase and the Cochrane Library Central Register up to May 2017 to identify published articles related to 5ARIs and the risk of MBC.2=0%, P=0.49). Estimates of total effects were generally consistent with the sensitivity.Summary effect estimates were calculated by a random-effect model, and tests for multivariable-unadjusted pooled risk ratios (RR) and heterogeneity, as well as the sensitivity analyses were conducted to assess publication bias. All four studies were conducted in a quality assessment according to the Newcastle Ottawa Scale system. The strength of association between 5ARIs and the prevalence of MBC was evaluated by using summarized unadjusted pooled RR with a 95% confidence interval [CI]. Four studies involving 595.776 participants, mean age range from 60 to 73.2 years old, were included in a meta-analysis, which produced a summary unadjusted RR of the risk of MBC for the treatment of 5ARIs of 1.16  and the multivariable-adjusted RR is 1.03, . There was no heterogeneity among included studies (IWe did not observe a positive association between the use of 5ARIs and MBC. The small number of breast cancer cases exposed to 5ARIs and the lack of an association in our study suggest that the development of breast cancer should not influence the prescribing of 5ARIs therapy. Five-alpha Reductase Inhibitors 5ARIs) finasteride and dutasteride are the competitive and specific inhibitors of 5α-reductase, an enzyme involved in the conversion of testosterone to 5α-dihydrotestosterone (DHT), which is the main androgen involved in the pathogenesis of benign prostatic hyperplasia (BPH) and androgenetic alopecia . These drugs act to prevent the transformation process of DHT to slow enlargement of the prostate, and to reduce the potentiation of AGA –4. FinasARIs finaFive-alpha Reductase Inhibitors have been investigated predominantly for their effects on BPH associated lower urinary tract symptoms (LUTS) and for the treatment of AGA on account of abundant 5ARI activities in the prostate and skin. These drugs have been widely used in clinical practice. There have been 50 worldwide case reports of male breast cancer (MBC) in BPH patients aged from 54 to 88 years (mean age 71 years old), who received 5 mg finasteride until 2009, including twenty-seven cases that occurred after finasteride treatment for a minimum of 1 year. There has been an increase in the number of case reports as pointed out in a recent report of Medicines and Health care products Regulatory Agency (MHRA). Patient information issued by the US Food and Drug Administration (FDA) for finasteride includes the following statements: “In rare cases, male breast cancer has been reported” and “The relationship between long-term use of finasteride and male breast neoplasia is currently unknown”  , studiesAll these information were recorded in a standardized form and the following data were sought from each study: year of publication, first author's name, ethnicity, diagnostic method, type of investigation, exact number of participants both in case and control groups, and relative risks as well. We assessed the quality of selected cohort studies and case-control studies according to the Newcastle-Ottawa quality assessment scale (NOS) ; the ins2 method . We carried out an additional analysis using the random-effects model described by DerSimonian et al.  , and then et al. , to obsen et al.  to deterThe diagram represents the flow of identification and inclusion of trials, as recommended by the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) statement  Figure-. In the The baseline characteristics of the included studies are shown in Two of the four studies which were evaluated by the Newcastle-Ottawa Scale (NOS) all are of the high level quality, ranging from 7 to 9. The results of quality assessment are summarized in 2=0%, P=0.49), which made us calculated with random effects model , and the multivariable-adjusted RR is 1.03, . There was no evidence of heterogeneity  to 1.21 , shown in In our assessment of publication bias, funnel plots showed balance, with points distributing around the verticals, indicating no obvious publication bias .To the best of our knowledge, this is the first meta-analysis which considers the effect of treatment of 5ARIs on the risk of MBC. From the selected articles, we clearly founded that MBC appears to be unrelated to the treatment of 5ARIs. This result was consistent with the study of Hagberg et al.  that colAs the disease incidence is less than 1% of female breast cancer, male breast cancer has not been studied well, and limited information is available regarding the epidemiology, pathogenesis, and treatment . We stilMBC is extremely rare, with an incidence in the general US population of <1%. It tends to be diagnosed at later stages than breast cancer in females, likely because of low awareness on the part of the patient and low suspicion by the physician. Clinical manifestations of MBC include breast mass (seen in 75% of patients), nipple retraction (9%), nipple discharge (6%), skin/nipple ulceration (6%) and Paget's disease of the nipple (1%) . Risk faMultifactorial effects are contributory to male breast tumorigenesis. We have no evidence to show that 5ARIs increased the risk of MBC according to our meta-analysis. We suggested that is because BRCA1 andBRCA2 mutations play a major role in the development of MBC.The BRCA1 and BRCA2 protein products are the key-point in DNA repair and cell cycle check point control , 44. TheThe strongest risk factor for MBC is the presence of an inherited BRCA2 mutation. The lifetime risk for breast cancer in a male BRCA2 mutation carrier is 7%, 80-100 times higher than for the general population –47. It iThe low incidence of MBC provides limited information on epidemiology, pathogenesis, and treatment to study. What is more, although we did not find any evidence of publication bias by testing funnel plots for obvious asymmetry, there still might be some unpublished studies that would nullify our results.The inadequate information and the relatively short times to onset in these cases make the causal association between MBC and finasteride unlikely. In the future, more epidemiological and clinical studies are required to further explore the association between 5ARIs and risk of MBC.


Recent several studies have showed that the nanog overexpression leads to poor prognosis in some kinds of cancer including hepatocellular carcinoma and gastrointestinal luminal cancer. However, the correlations between prognosis and clinic-pathological features and nanog overexpression in lung cancer are still not well-known. Thus, we performed a meta-analysis to evaluate the role of nanog in lung cancer. An electronic retrieval for related studies was conducted in PubMed, Cochrane Library, Web of Science, EMBASE databases, Chinese CNKI, and the Chinese Wan Fang database up to May 2018. The relationships between nanog overexpression and overall survival (OS) and disease-free survival (DFS) as well as clinic-pathological features in lung cancer were investigated. Pooled hazard ratios (HRs) and odds ratios (ORs) with 95% confidence intervals (CIs) were calculated by STATA12. P ≤ 0.001) and DFS . Moreover, overexpression of nanog was significantly related to differentiation , lymph node metastasis  and tumor size , and no correlation with T stage, TNM, stage, and gender. 11 studies containing 1422 patients were identified in our meta-analysis. The overexpression of nanog showed decreased OS (HR = 1.83, 95% CI = 1.49-2.25, Our results suggested that nanog overexpression, a hazard factor of differentiation, lymph node metastasis, and tumor size, may predicate decreased OS and DFS for lung cancer. Lung cancer is ranked at the first in both incidence and mortality worldwide , 2. SignAt present, the latest theory shows that tumors are a kind of stem cell disease. Malignant tumors include a small portion of cancer stem cells (CSCs) and act as a pivotal part in the formation and growth of tumors. These CSCs have the ability to multiply, self-renew and differentiate, and express similar molecular markers and gene products . Nanog wTo deeply understanding the relation between nanog overexpression and prognosis in lung cancer, we performed the present meta-analysis to assess the influence of nanog on survival and clinic-pathological parameters in patients.The electronic database including PubMed, Cochrane Library, Web of Science, EMBASE database, Chinese CNKI, and the Chinese Wan Fang database was searched (Last update May 2018). Searches contained the terms “nanog or NANOG” (abstract/title) and “cancer or tumor or carcinoma or neoplasm” (abstract/title) and “lung or pulmonary” (abstract/title). All qualified studies were acquired and all references for those selected articles were screened and evaluated. Some review articles were manually retrieved to look for other qualified studies which were then assessed for inclusion by two reviewers (Wei Cheng and Juanjuan yuan). Divergences were resolved by consultation.The included criteria for eligible studies in our meta-analysis are (1) patients were diagnosed by histopathologic examinations; (2) the expression of nanog was assayed by immunohistochemistry (IHC) or reverse transcription polymerase chain reaction (RT-PCR); (3) the results contained the survival information (including OS or DFS) or clinic-pathological variables; (4) hazard ratios for OS or DFS can be acquired from studies or calculated from the survival curves.The data was extracted from all qualified studies by two authors (Wei Cheng and Juanjuan Yuan). All the extracted data contained first author, publication year, country, number of cases, detect methods of nanog, antibody used, cut-off value of nanog, hazard ratios (HR) and 95% confidence intervals (CIs) , and somThe studies quality was evaluated by Newcastle-Ottawa-Scale (NOS) criteria . The stu2 test and Q test. We considered heterogeneity present when P ≤ 0.1 for the Q test or I2 > 50%. In the absence of statistical difference for heterogeneity, we used the Mantel-Haenszel method in the fixed-effect model for our meta-analysis. Otherwise, the DerSimonian and Laird method in the random-effect model was chose. The effect of nanog overexpression on survival and clinic-pathological features was evaluated with pooled HRs and ORs with 95% CIs. Publication bias was assessed by Begg's funnel plot and Egger's test [ P < 0.05 was considered as statistical difference [We divided the original aims into two categories for this meta-analysis. The first aim was to estimate the prognostic value of nanog overexpression on OS and DFS. HRs and 95% CIs were calculated by the survival data extracted from Kaplan-Meier curve with Engauge Digitizer version 4.1 as described before when not directly gained in studies –28. The r's test , 30. P <fference . The sou619 potential relevant studies were searched from the databases according to the search terms. As shown in P = 0.029, I2 = 51.6%,  P = 0.004, I2 = 77.2%,  P ≤ 0.001,  P = 0.006, Heterogeneity was significant among the included studies when assessing relationship between nanog overexpression and OS and DFS for lung cancer . Our outcome also indicated that high nanog expression was associated with poor OS and DFS compared with low nanog expression in adenocarcinoma subgroup . Finally, our subgroup analysis showed that nanog overexpression was always linked to shorter OS and DFS in lung cancer.Subgroup analysis was used to explore possible sources of heterogeneity among OS  and DFS  P = 0.655 and Egger's test of P = 0.656; Begg's funnel plot and Egger's test were performed to evaluate the publication bias for OS and DFS in lung cancer patients in included studies. As shown in Figures In order to appraise the effect of single study on the pooled HRs in OS and DFS because of significant heterogeneity, we carried out sensitivity analysis by estimating the average HRs in the absence of each study. The results demonstrated that our meta-analysis was statistically reliable Figures .Lung cancer is a malignant tumor with high morbidity and mortality. Although there are many treatment strategies for lung cancer, the therapeutic effectiveness is still not satisfactory. It is emergency for us to explore the new mechanism of metastasis and recurrence and look for related prognostic markers and targets of therapeutic interventions to improve the prognosis for lung cancer.More and more studies have showed that there is a small count of cells with self-renewal and differentiation in tumors. Their characteristics are similar to normal stem cells. We call them CSCs. Increased CSCs are often associated with tumor progression, relapse, and drug resistance , 33. TheOur outcomes showed that nanog overexpression have been involved with poor OS and DFS in lung cancer. For clinic-pathological features involved in lung cancer, we found that nanog overexpression was associated with differentiation, lymph node metastasis, and tumor size. The reasons for this may be that nanog can promote invasion, metastasis, and cell proliferation in lung cancer. Our results also demonstrated that no obvious relation was found between nanog overexpression and T stage, TNM stage, and gender. Because of obvious heterogeneity in OS and DFS, we carried out subgroup analysis based on pathological types, publication year, NOS score, and country. Our subgroup results showed that publication year might be considered as a source of heterogeneity. There was no significant heterogeneity in both publication year subgroup (2010-2016) and publication year subgroup (2017) for OS and DFS. Our results indicated that nanog overexpression led to poor OS and DFS in two subgroups. Du to absence heterogeneity for OS and DFS in adenocarcinoma subgroup based on pathological types, our outcome also demonstrated that there was a decreased OS and DFS for lung adenocarcinoma patients with high nanog expression compared with low nanog expression. We performed sensitivity analyses by evaluating the average HRs in the absence of each study. The results indicated that our meta-analysis was statistically reliable. And currently, the heterogeneity cannot be well elaborated and still requires some high quality studies with large sample. In short, our results suggested that nanog overexpression may indicate poor OS and DFS and susceptibility to poor differentiation and undifferentiation, lymph node metastasis, and tumor size (≥3cm). So, it is possible for us to improve OS and DFS for patients with lung cancer by targeting nanog therapy in future. We can also consider determining treatment strategies according to nanog expression level.Publication bias is an important limitation in meta-analysis, because some studies with negative results are more difficult to be accepted for publication. Thus we should encourage some researchers to publish their studies including some negative results. Our results demonstrated that no significant evidence of publication bias was found in our included studies.There are still so many other limitations in our study. First of all, prognosis data extracted from survival curves might be less reliable than reported directly in studies. Second, the antibody used, IHC cell-scoring method, and the cut-off value were not defined similarly in partial studies. Third, the heterogeneity of the OS and DFS is significant, although we performed subgroup analysis and sensitivity analyses. These factors may contribute to potential publication bias. Our sensitivity analyses indicated that the results were stable and the heterogeneity did not influence the analysis results.To summarize, our results demonstrated that nanog overexpression, a hazard factor of differentiation, lymph node metastasis, and tumor size, may contribute to poor OS and DFS for lung cancer. Nanog might be a bad prognostic marker for lung cancer.


Longitudinal, well phenotyped, population-based cohort studies offer unique research opportunities in the context of genome-wide association studies (GWAS), including GWAS for new-onset (incident) cardiovascular disease (CVD) events, the assessment of gene x lifestyle interactions, and evaluating the incremental predictive utility of genetic information in apparently healthy individuals. Furthermore, comprehensively phenotyped community-dwelling samples have contributed to GWAS of numerous traits that reflect normal organ function  and for many traits along the CVD continuum . These GWAS have heretofore identified many genetic loci implicated in normal organ function and different stages of the CVD continuum. Finally, population-based cohort studies have made important contributions to Mendelian Randomization analyses, a statistical approach that uses genetic information to assess observed associations between cardiovascular traits and clinical CVD outcomes for potential causality. As a brief introduction, we would like to highlight important design features of population-based studies. As opposed to hospital-based referral samples, population-based epidemiological studies examine community-dwelling or random samples from the general population. As such, study participants are not selected based on a given disease, but rather to represent the general population of the areas sampled, so that observations from such a sample are generalizable to the underlying source population. It has to be kept in mind, though, that the response rate of some landmark cohort studies is rather low , which Third, many population-based cohort studies perform Since many community-dwelling samples are representative of the general population, population-based studies have served as reference (“control”) samples for many genetic case-control analyses. In essence, genetic case-control studies compare allelic frequencies of genetic variants in prevalent cases (patients who have the disease of interest when they are sampled) and controls. Ideally, the control sample captures the distribution of the exposure  in the source population from which the cases were derived . TherefoThe broad and highly standardized phenotyping of their study participants has allowed many different contributions of population-based cohort studies to GWAS. Specifically, researchers from population-based studies have performed and contributed to numerous GWAS for traits along the cardiovascular disease continuum, including traditional CVD risk factors [e.g., lipids , blood pFurthermore, given the large proportion of apparently healthy individuals in population-based cohort studies , these studies conducted GWAS of many traits that reflect relatively normal organ function, including biomarkers of cardiac structure and systolic and diastolic function . These sIt is an important and growing area of research to quantify the contribution of genes and of different lifestyle factors (and their interactions) to inter-individual variation in cardiovascular risk factor levels and disease risk. Since well phenotyped cohort studies usually have comprehensive genetic data and detailed lifestyle information available, population-based studies represent an ideal setting to study gene x lifestyle interactions. The interaction of a genetic risk score (based on 50 SNPs) and a lifestyle score  on the incidence of CAD has been analyzed in several large community-based cohorts . Key obsThe same lifestyle score as in Reference  was usedPCSK9 gene and a PCSK9 genetic score were associated with lower circulating LDL cholesterol levels and reduced risk of cardiovascular events in the general population  mutations in the PCSK9 (proprotein convertase subtilisin/kexin type 9) serine protease gene were initially identified in families with autosomal dominant hypercholesterolemia . Subsequpulation . Recentlpulation .Due to the availability of repeated measures over time, cohort studies are also suitable to explore the genetic underpinning of changes in cardiovascular risk factors over time, and of the progression of subclinical CVD traits longitudinally. For example, a GWAS for carotid IMT measured at different time points over a 10-year period has recently been published . FurtherOn a parallel note, a genetic risk score consisting of 29 SNPs was not only associated with blood pressure and hypertension prevalence at baseline, but also with new-onset hypertension and change in blood pressure over the life course in a large Swedish cohort study .The longitudinal character of population-based cohort studies allows genetic variation to be studied in relation to disease incidence. For example, population-based cohort studies have facilitated GWAS for incident heart failure , incidenIt is well known that analyses based on prevalent disease cases and those based on incident cases might reveal different results if the association between an exposure and the disease outcome differs by disease severity or disease duration  . In ordeFurthermore, community-based prospective cohorts allow assessing whether genetic information improves risk prediction models beyond traditional risk factors. It was, indeed, one of the main motivations of the human genome project to use genetic information to predict disease risks in healthy individuals and to predict the response to a given therapy among patients. Several analyses conducted in various population-based cohorts assessed whether genetic variation – e.g., in an aggregated form as risk scores – improved performance measures of risk prediction models for a first CVD event, including discrimination, calibration, and reclassification . AlthougGenetic information in population-based cohort studies has also been used to assess causality between cardiovascular risk factors or circulating biomarkers and cardiovascular outcomes (incident CVD events) using instrumental variable analyses, a statistical approach referred to as Mendelian Randomization (MR) . This tePopulation-based studies have substantially improved our understanding of the genetic architecture of normal and abnormal organ function, CVD risk factors, circulating biomarkers, subclinical disease, and overt CVD traits over the life course. Furthermore, they were essential in exploring gene x lifestyle interactions and in evaluating genetic variation in the context of risk prediction models for incident CVD. In addition, population-based cohort studies provided great opportunities to conduct GWAS for incident CVD events, such as MI, stroke and heart failure, and thereby, to overcome classic limitations of case-control GWAS including prevalence-incidence bias. Finally, population-based cohort studies used genetic information as instrumental variables to assess whether cardiovascular risk factors or biomarkers are causally related to clinical CVD .WL and RV wrote the article together.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 relative risk of dementia per 1 cup/day increment of coffee consumption was 1.01  0.98–1.05; p = 0.37). Meta-analysis of five studies that focused on Alzheimer’s disease revealed no association between coffee consumption and Alzheimer’s disease and no deviations from a linear trend (p = 0.79). The relative risk of Alzheimer’s disease per 1 cup/day increment of coffee consumption was 1.01 . These results do not support an association between coffee consumption and an increased risk of overall dementia or Alzheimer’s disease specifically, but further research on the association of coffee consumption with dementia risk is needed.Coffee consumption is associated with a reduced risk of several diseases but uncertainty remains about the influence of coffee consumption on the risk of dementia. We performed a dose-response meta-analysis to summarize the prospective data on coffee consumption and associated risk of dementia and Alzheimer’s disease. We identified studies by searching PubMed (from January 1966) and Web of Science (from January 1945) through 4 October 2018 and by scrutinizing the reference lists of pertinent publications. Two researchers independently reviewed the literature. Results were combined using a restricted cubic spline random-effects dose-response meta-analysis based on a one-stage approach. Eight relevant prospective studies were identified. These studies included 7486 dementia cases diagnosed among 328,885 individuals during an average follow-up of 4.9–25 years. Meta-analysis of all eight studies indicated no statistically significant association between coffee consumption and the risk of dementia and no deviations from a linear trend ( Dementia is an increasing global health concern due to the aging population. Globally, about 50 million individuals had dementia in 2015 and this number is projected to increase to 132 million by 2050 [Evidence from observational studies indicates that coffee consumption may lower the risk of several diseases, including diabetes, stroke, coronary heart disease, certain cancers, Parkinson’s disease and gout ,4,5,6,7.To further clarify whether coffee drinking is related to the risk of dementia or AD, we performed a dose-response meta-analysis of prospective studies. This meta-analysis includes data from three additional prospective studies ,18,19, wThe PRISMA guidelines were followed for this meta-analysis. We searched PubMed from 1 January 1966 through 4 October 2018, using the search terms “coffee” (All Fields) and “dementia” (All Fields) or “coffee” (All Fields) and “Alzheimer’s disease” (All Fields). Moreover, Web of Science was searched from 1 January 1945 through 4 October 2018 using the search query “TOPIC: (coffee) and TOPIC: (dementia) or TOPIC: (coffee) and TOPIC: (Alzheimer’s disease)”. We searched for additional studies by scrutinizing the reference lists of relevant articles. Two researchers (S.C.L. and N.O.) independently reviewed the literature. A summary of the literature search, screening and study selection is presented in The eligibility criteria in this meta-analysis included prospective studies that reported relative risk (RR) estimates with corresponding 95% confidence intervals (CI), or sufficient data to estimate these, of AD or all-cause dementia for at least three categories of coffee consumption. Cross-sectional, case-control and Mendelian randomization studies as well as studies reporting results on cognitive impairment only were excluded. From each study, the authors (S.C.L. and N.O.) extracted the following information: study name and country in which the study was performed, sample size, number of cases, how cases were ascertained, age of participants, proportion of men, duration of follow-up, potential risk factors adjusted for in the multivariable analysis, the RRs with their 95% CI for each exposure category and the number of cases and total number of participants or person-years in each category. We extracted the multivariable RRs that were adjusted for the largest number of possible confounders. The Newcastle-Ottawa Scale was used to assess study quality . The assI2 statistic [Our primary analysis included dementia as the outcome  according to the quantitative categories of coffee consumption. Studies that reported results on AD only were included in the primary analysis but were excluded in a sensitivity analysis. In addition, we performed analyses of coffee consumption in relation to AD specifically. We performed a random-effects dose-response meta-analysis using a one-stage approach and restricted maximum likelihood estimation . We assetatistic . Publicatatistic . The staEight prospective studies that examined the association of coffee consumption with the risk of AD and/or all-cause dementia were eligible for inclusion in the current meta-analysis ,30,31,32p = 0.08) (p = 0.37). There was no heterogeneity between the results of individual studies  and no indication of publication bias (p = 0.48) (funnel plot is presented in p = 0.58) per 1 cup/day increment of coffee consumption.Meta-analysis combining the data from the eight studies revealed no statistically significant association between coffee consumption and the risk of dementia and no deviations from a linear trend ( = 0.08) . The RRs = 0.08) . The RR p = 0.79) (p = 0.80), with no statistically significant between-study heterogeneity .Meta-analysis of the five studies with results on AD revealed no association between coffee consumption and the risk of AD and there was no indication of departure from linearity ( = 0.79) . The RR This contemporary meta-analysis of eight prospective studies provides no evidence that coffee consumption is associated with the risk of AD or overall dementia. However, we cannot rule out that we may have overlooked a weak association.Three studies reporting results on coffee consumption and AD risk were non-eligible for inclusion in this dose-response meta-analysis because results were not reported in different quantitative categories of coffee consumption ,21,22. TA strength of this meta-analysis based on the data from eight prospective studies is that the association between coffee consumption and the risk of dementia could be assessed with relatively high precision. Furthermore, we were able to investigate the dose-response relationship between coffee consumption and dementia risk. A limitation of this meta-analysis is that self-reported coffee consumption is subject to measurement errors, which may have resulted in attenuated results. The exact amount and concentration of coffee consumed are difficult to measure by a questionnaire and therefore, this meta-analysis may not have captured the true dose-response relationship between coffee consumption and dementia risk. Likewise, the diagnosis of dementia and AD differed among studies and some degree of misclassification of dementia is likely to have occurred in most studies. Another shortcoming is that the findings were based on data from observational studies and thus, residual confounding and reverse causality cannot be ruled out as explanations for the lack of an observed association of coffee consumption with dementia risk. Finally, we were unable to examine whether the intake of caffeinated or decaffeinated coffee specifically is associated with the risk of dementia. In conclusion, the available data from observational prospective studies provide no evidence that coffee consumption is associated with the risk of AD or all-cause dementia. Nevertheless, there have only been a few studies examining the relationship of coffee consumption with the risk of dementia and AD specifically are relatively few, while the misclassification of coffee consumption and of dementia and may have attenuated the results. Therefore, additional studies that are well conducted are necessary to establish whether coffee consumption influences the risk of dementia or AD.


Colorectal cancer (CRC) is highly prevalent cancer, which should be genetically studied among different peoples of the worldXRCC3T241M, XRCC3A17893G and, for the first time, XRCC7 I3434T polymorphisms on CRC risk in Khorasan Razavi Province, Iran. The aim of this study was to evaluate the effect of In this case-control study, 180 patients with CRC and 160 sex- and age-matched healthy controls were studied. Genotypes were determined by RFLP-PCR and ARMS-PCR. XRCC3 C/T genotype than in the CC genotype . In the case of the XRCC7 I3434T polymorphism, CRC risk was significantly (4.3 fold) higher in I/T+T/T variant subjects compared to the I/I genotype . Moreover, the XRCC3 A17893G polymorphism did not correlate with CRC. In addition, there was no significant difference between the distribution of genotypes of the three studied polymorphisms with demographic and clinicopathological features in the CRC patients. The incidence of CRC was observed to be significantly more in a heterozygous XRRC3 and XRCC7 genes are involved in CRC and should be considered as a risk factor.Polymorphisms of  Colorectal cancer (CRC) is a prevalent cancer worldwide, especially in developed countries, and nowadays, it is increasing in developing countries such as Iran . Therefore, CRC should be studied in terms of diagnosis, prevention and treatment. XRCC3 and DNA-dependent protein kinase (DNA-PK), contribute to the repair of damaged DNA . If repair factors themselves have problems, repair will likely be confronted with problems and thus mutations and then cancer may occur.Human beings are exposed to a large number of exogenous and endogenous agents that damage genes. Harming agents include various types of air pollution, some types of food and diet, Tobacco, Alcohol, ionizing radiation, ROS and several other factors . If damages to a gene are not repaired, they will be mutated in the next generation of cells . A large number of such mutations can lead to tumors and cancers, such as CRC. Depending on the type of damage, there are several pathways for restoration of genes. The main pathways to repair DNA are excision repair (ER) and double-strand break repair (DSBR). DSBR consists of two types of mechanisms, including homologous recombination (HR) and non-homologous end joining (NHEJ) . In each mechanism, various proteins and enzymes, including As we know, after completing the human genome project, it became clear that our genes have a lot of polymorphisms. One of the most abundant types of polymorphisms is single nucleotide polymorphism (SNP). The existence of a SNP in the exon region of a gene may lead to a change in one of the amino acids in the protein. By replacing an amino acid in a protein, the conformation and function of that protein may be altered and it cannot act properly. If SNPs are present in non-exon regions of a gene, they may affect regulation of transcription or mRNA processing and turnover . Thus, in general, polymorphisms may play an important role in the development of a cancer. XRCC3) and X-ray repair complementing defective repair in Chinese hamster cells 7 (XRCC7) are two genes that were studied in the present work. The protein encoded by the XRCC3 gene is XRCC3 and the product of the XRCC7 gene is DNA-PKcs. Both of the proteins are involved in the DSBR mechanism, i.e. XRCC3 in HR and DNA-PK in NHEJ . A common SNP in exon 7 of the XRCC3 gene leads to replacement of an amino acid at position 241 (Thr241Met) of the protein and the IVS5–14  polymorphism is in intron 5 of the gene . So far, several studies have been conducted on these SNPs in various cancers and in different human races. Some of these studies have proven the association between these polymorphisms and cancer while some others have found no relationship between them, such as studies on colorectal adenoma , lung  and breast cancer . Further, there are scarce studies on the association between the common genetic Ile3434Thr polymorphism (rs7830743) XRCC7 and cancers. Moreover, few studies conducted on XRCC3 have also reported controversial results, with some studies confirming the relationship between XRCC3 and cancer and some others rejecting it .X-ray repair complementing defective repair in Chinese hamster cells 3 . In this case-control study, approximately 5 ml of peripheral whole blood were collected from each of the 180 cases of sporadic CRC and 160 healthy individuals, as controls, in EDTA containing tubes and stored at -80°C until analyzed. The mean (±SD) age of the patients was 57.9±14.4 years and they included 77 (42.8%) females and 103 (57.2%) males. The control group, including 70 (43.8%) females and 90 (56.2%) males, had an average (±SD) age of 57.2±13.9 years. The patients were randomly selected from CRC patients referred to the Reza Radiotherapy and Oncology Center in Mashhad and the 22 Bahman Hospital in Neyshabur. Healthy individuals were selected from among volunteers whose age and gender were matched to the patients. Before collecting blood samples and filling out questionnaires, information was provided to the patients and the healthy individuals and then written consent forms were obtained from all the participants. The clinicopathological information of the patients was collected by the manual review of their pathology reports and hospital records.DNA Extraction and GenotypingXRCC3 gene and by an amplification refractory mutation system-PCR called the ARMS-PCR method in the XRCC7 polymorphism, using forward (F) and reverse (R) primers listed in Kits required to isolate genomic DNA were purchased from the Korean company, Bioneer. Thus, DNA extraction was carried out according to the manufacturer’s instructions and stored at –20˚C until used for genotyping. Genotyping of two SNPs was carried out by restriction fragment length polymorphism-PCR (RFLP-PCR) in the 2, 100 ng DNA, 12.5 pmol of each primer, and 1 U Taq DNA polymerase. The 358bp amplified product for XRCC3 (rs861539) was digested with FatI. The wild-type allele Thr was identified by the presence of a 358bp band (indicative of the absence of the FatI cutting site), while the mutant allele Met was detected by 200 and 158bp bands (XRCC3 (rs1799796), a 430bp amplified product was digested with AluI. The wild-type allele G was identified by the presence of a 430bp band (indicative of the absence of the AluI cutting site), while the mutant allele A was detected with the appearance of 226bp and 204bp bands  and common reverse (CR) primers, while 116bp and 165bp allele-specific amplicons were amplified using CF-SC and CR-ST primer pairs, respectively . Then, the related p-value was calculated, which is available at the https://www.socscistatistics.com/pvalues/chidistribution.aspx. Moreover, the genotype and allele frequencies were compared in different groups using the chi-square test and the Fisher’s exact test (two-tailed). The odds ratios (OR) and 95% confidence intervals (CI) were calculated by means of a logistic regression model. SPSS 20.0  was used to perform statistical calculations. All p-values less than 0.05 were considered significant.After doing laboratory work, collecting demographic and clinical data and performing statistical analysis, our findings are summarized in Regarding the p-values in XRCC3 Thr241Met and XRCC7 Ile3434Thr were involved in the development of CRC, but the XRCC3 rs1799796 polymorphism did not correlate with this cancer. Accordingly, the incidence of CRC was significantly more in the heterozygous XRCC3 241Thr/Met (CT) genotype than in the CC genotype . Generally, individuals with the T allele are more likely to have CRC . In the case of the XRCC7 Ile3434Thr polymorphism, CRC risk was significantly (4.3 fold) higher in AG+GG variant subjects compared to the AA genotype . Therefore, the CT marker is co-dominant and the C allele is not dominant or recessive. In general, those who simultaneously had T and C alleles were more likely to have colorectal cancer . In the case of the Ile3434Thr XRCC7 polymorphism, the risk of CRC in individuals with a heterozygote AG (Ile/Thr) genotype was 2.2 times higher than those with normal AA homozygote, although the p-value was greater than 0.05. Also, the risk of CRC was significantly 4.3 fold greater in AG+GG variant subjects compared to the AA genotype . Thus the AG is a co-dominant marker.According to other results of the present study, which are presented in XRCC3 Thr241Met, XRCC3 rs1799796 and XRCC7 Ile3434Thr genotypes in the control group were in the Hardy-Weinberg equilibrium, as demonstrated by the lack of any significant difference between their observed and expected frequencies . Therefore, this suggests that the control subjects may represent the general population. The XRCC3 and XRCC7. If any of such processes has problems, the repair will not be performed correctly and will appear as a mutation in a new cell.There are several processes to repair DNA damages, and in each mechanism, there are a large number of proteins, including XRCC3Thr241Met, XRCC3rs1799796 and XRCC7 Ile3434Thr polymorphisms were different in CRC and healthy individuals as they are susceptibility factors for the development of CRC. The present study was conducted to determine whether the XRCC3 protein is one of the RAD51 parallogues and participates in HR through interaction with other proteins. There are studies that have shown that XRCC3 deficiency may interfere with DSBR . It is therefore assumed that this protein plays an important role in the DNA repair. The presence of the rs1799796 polymorphism in the intron region of the XRCC3 gene, if reduces gene expression, may interfere with DNA repair and contribute to the abundance of mutations and cancer risk. However, according to the results of our study, it was realized that this polymorphism did not affect the regulation of XRCC3 gene expression. Moreover, the findings of the present study revealed that there was no significant difference in the distribution of the frequency of the XRCC3rs1799796 polymorphism between the CRC and control groups . It should be noted that a number of studies have similar results to those of our study. For example, a study found no association between this polymorphism and colorectal adenoma in the United States  with breast cancer in Saudi Arabia and Taiwan . However, a significant relationship was observed between the XRCC3rs1799796 polymorphism and ovarian cancer risk in a study conducted in China .In humans, the XRCC3Thr241Met polymorphism, several investigations observed that the presence of Met with a hydrophobic methyl sulfur group instead of Thr with a hydrophilic hydroxyl group at the position 241 of the XRCC4 protein was significantly related to high DNA adduct levels . Scientists have determined that XRCC3 has only a domain with functional activity and the 241Thr/Met variant is located in this domain. Therefore, it is believed that the XRCC3 (241Met) variant is involved in susceptibility to cancer . The present study confirmed that the XRCC3Thr241Met polymorphism had a significant relationship with CRC risk . The increased risk of cancer associated with this polymorphism has also been proven in several other studies on CRC , gastric cancer , ovarian cancer , endometrial cancer , and prostate cancer . However, some studies have found no significant relationship between this polymorphism and cancers, such as breast cancer  and hepatocellular carcinoma .With regard to the XRCC7 gene . There is a polymorphism in the XRCC7 gene, which results in the replacement of the hydrophobic isoleucine with the hydrophilic threonine at position 3434 in the FRAP-ATM-TRRAP domain of the DNA-PK protein. Therefore, it can be assumed that this substitution may interfere with the functioning of the protein and thus the DNA repair will not be done properly, which ultimately leads to cancer. However, so far, little research has been carried out on the association between the XRCC7 Ile3434Thr polymorphism and cancers. According to our knowledge, the present study was the first investigation on CRC, and the findings indicated that the 3,434Thr/Ile variant contributed to this cancer . Previously, a study carried out in Iran showed that this polymorphism was also related to thyroid cancer . Therefore, the results of these two studies indicated that the XRCC7 Ile3434Thr polymorphism could be a tumor marker in Iran. However, a study conducted in China showed that the XRCC7 Ile3434Thr polymorphism did not correlate with cancer .Another protein involved in the DSB repair is the DNA-PK protein derived from the All conflicting relationships mentioned in various studies are expected in different countries and regions, which may be related to ethnic difference, sample size and various other variables.The interaction between gene polymorphism with demographic factors and pathological symptoms of CRC was investigated in a limited number of studies, including the current study. However, no significant relationship was found between them in almost all of these studies, which is in line with our findings . For example, a study about ornithine decarboxylase gene polymorphism (+316 ODC1 rs2302615) showed that age, gender, stage, colon/rectum site, tumor grade, and histologic subtype were not related to genotypes and alleles, with all p-values>0.05 . Another study found that the interaction between transforming growth factor-β1 gene promoter -509C/T polymorphism and CRC was only due to the relationship between gender and stage with CC genotype, but not due to other noted variables . The existence of the CC genotype in men has been associated with cancer development and progression.XRCC3Thr241Met polymorphism, but not its rs1799796 polymorphism, plays a role in the development of CRC. Moreover, the XRCC7 Ile3434Thr polymorphism plays a significant role in increasing CRC risk. However, since it has been less studied, it is suggested to further examine its effect on more samples. According to the findings of the present study, the studied polymorphisms in XRCC3 and XRCC7 did not cause any progression, metastasis or other parameters in CRC.In general, it can be concluded that the


Background and Objectives: Several studies inspected the impact of P2X7 polymorphisms on individual susceptibility to tuberculosis (TB), but the findings are still controversial and inconclusive. To achieve a more precise estimation, we conducted a meta-analysis of all eligible studies on the association between P2X7 polymorphisms and TB risk. Materials andMethods: Relevant studies were identified by searching the PubMed, Web of Science, Scopus and Google scholar databases up to November 2018. Twenty-four full-text articles were included in our meta-analysis. The strength of association between P2X7 polymorphisms and TB risk was evaluated by odds ratios (ORs) and 95% confidence intervals (95% CIs) under five genetic models. Results: The findings of this meta-analysis revealed that the rs3751143 variant significantly increased the risk of TB in heterozygous codominant , homozygous codominant , dominant , recessive , and allele  genetic models. Stratified analysis showed that rs3751143 increased the risk of pulmonary tuberculosis (PTB) and extrapulmonary tuberculosis (EPTB) in all genetic models. Furthermore, the rs3751143 increased risk of TB in the Asian population. The findings did not support an association between the rs2393799, rs1718119, rs208294, rs7958311, and rs2230911 polymorphisms of P2X7 and TB risk. Conclusions: The findings of this meta-analysis suggest that P2X7 rs3751143 polymorphism may play a role in susceptibility to TB in the Asian population. More well-designed studies are required to elucidate the exact role of P2X7 polymorphisms on TB development. Mycobacterium tuberculosis (MTB). It remains a serious public global health problem. According to the World Health Organization (WHO) report, there were an estimated 10.4 million new cases of TB worldwide and approximately 1.3 million deaths in 2016 [Tuberculosis (TB) is a chronic infectious disease caused by the bacillus  in 2016 ,6.P2X7 gene is mapped to chromosome 12 (12q24.31). It encodes cell-surface nucleotide receptors called P2X7 receptor (P2X7R) [2+ and Na+ and the efflux of K+ [The human  (P2X7R) . Activatux of K+ . In M. tux of K+ ,11.P2X7R gene is indeed polymorphic. The precise correlation between the P2X7 polymorphisms and susceptibility to TB is not completely documented. Several single nucleotide polymorphisms (SNPs) have been revealed that affect the function of this receptor which cause P2X7R loss-of-function (LOF) or gain-of-function (GOF) [P2X7 polymorphisms and risk of tuberculosis in various populations, but the findings were inconsistent [P2X7 polymorphisms and tuberculosis risk.The on (GOF) . Many stnsistent ,33,34,35P2X7 polymorphisms and TB risk were searched. The search key words used were “P2X7 or P2X7R” and “tuberculosis” and “polymorphism or variant”. P2X7 polymorphisms and TB risk; the frequencies distribution of alleles and genotypes in patients and controls can be extracted. The exclusion criteria were studies that are not associated with P2X7 polymorphisms and TB risk; overlapping data, conference papers, reviews, meta-analyses; no sufficient data reported.The PubMed, Web of Science, Scopus, and Google scholar databases for all potentially eligible research articles up to November 2018 on the relationship between Two investigators independently inspected and evaluated the articles for eligibility according to inclusion and exclusion criteria. The following data were recorded from the selected studies such as the first author’s name, publication year, ethnicity, genotyping methods, genotype and allelic profile, as well as the source of controls.P2X7 polymorphisms and TB susceptibility. The significance of the pooled OR was determined by the Z-test, and a p-value less than 0.05 was considered significant. Heterogeneity between the studies was estimated by Q statistic and the I2 test. p < 0.10 designated significant heterogeneity. If heterogeneity did not exist, a fixed-effects model was used to calculate the pooled ORs; otherwise, a random-effects model was utilized.The chi-square test was used to check whether genotypes within the controls conformed to the Hardy-Weinberg equilibrium (HWE). We calculated the pooled odds ratios (ORs) and corresponding 95% confidence intervals (CIs) to assess the association between the p value < 0.05 was considered a significant publication bias. Sensitivity analysis was done by neglecting each study in turn to assess the quality and consistency of the results. All statistical analyses were executed using STATA v14.1 software .Publication bias was inspected with the funnel plot and an asymmetric plot suggests a possible publication bias. Funnel plot asymmetry was further measured using Egger’s linear regression test. Through a comprehensive literature search and selection based on the inclusion criteria, 59 relevant case-control studies from 24 selected articles ,33,34,35p = 0.0006, AC vs. AA), homozygous codominant , dominant , recessive , and allele  genetic models  polymorphism significantly increased the risk of overall TB. Stratified analysis of this polymorphism significantly increased the risk of PTB and EPTB. Also, the rs3751143 polymorphism increased the risk of TB in the Asian population. Findings did not support an association between rs2393799 (−762 C > T), rs1718119 (Thr348Ala), rs208294 (His155Tyr), rs7958311 (Arg270His), and rs2230911(Thr357Ser) polymorphisms and TB risk.Mounting evidence proposed that host genetic factors are implicated in tuberculosis susceptibility ,6. The Pon to TB ,33,34,35n = 10) on the association between P2X7 rs3751143 polymorphism and PTB risk and found that this variant significantly increased the risk of PTB. Another meta-analysis (n = 11) performed by Alshammari et al. [Ge et al.  performei et al. . Anotheri et al.  revealedi et al.  on the aP2X7 is rs3751143  polymorphism located in exon 13, accountable for LOF. This polymorphism affects the sensitivity of P2X7R to ATP and may contribute to increased susceptibility to MTB infection in humans [P2X7 and decrease the expression of P2X7R. The relationship between the rs2393799 polymorphism and susceptibility to TB is still debated [Several polymorphisms have been described that cause P2X7R loss-of-function (LOF) or gain-of-function (GOF) . The comn humans . The fin debated ,32,33,34 debated . Limited debated ,34. PoolUp until now, only 3 studies investigated the association between rs1718119  polymorphism and TB risk ,31,34. OPorphyromonas gingivalis, a bacterial carcinogen, plays a key role in cancer development by inhibiting apoptosis through several mechanisms. It has been shown that this bacterium secretes an anti-apoptotic enzyme nucleoside diphosphate kinase (NDK) which cleaves ATP and prevents proapoptotic P2X7 receptor activation, consequently modulating ATP/P2X7-signaling pathway [ pathway . It has  pathway .Some limitations of our meta-analysis should be acknowledged. Firstly, heterogeneity between studies was evident, which might distort the conclusion of this meta-analysis. Heterogeneity may be partly arising in the differences of ethnicities. Secondly, the sample sizes for some polymorphisms were small. Therefore, the results of this meta-analysis should be interpreted with caution.P2X7.Despite these limitations, however, there are still some advantages to having done this meta-analysis. First, this meta-analysis involved more studies than the previous meta-analyses, so the statistical power of our study is higher than the published meta-analysis. Second, we evaluated six polymorphisms of P2X7 rs3751143 polymorphism may serve as a risk factor for TB in the Asian population. However, further well-designed studies with large sample sizes are necessary to confirm our findings.Overall, our meta-analysis proposed that


MSX1 and STX18, has been associated with atrial septal defects (ASD) in multiple European and Chinese cohorts. Here, genotyping data from the UK Biobank was used to test for associations between this locus and congenital heart disease in adult survivors of left ventricular outflow tract obstruction (n = 164) and ASD (n = 223), with a control sample of 332,788 individuals, and a meta-analysis of the new and existing ASD data was performed. The results show an association between the previously reported markers at 4p16 and risk for either ASD or left ventricular outflow tract obstruction, with effect sizes similar to the published data . Differences in allele frequencies remained constant through the studied age range (40–70 years), indicating that the variants themselves do not drive lethal genetic defects. Meta-analysis shows an OR of 1.35  for the association with ASD. The findings show that the genetic associations with ASD can be generalized to adult survivors of both ASD and left ventricular lesions. Although the 4p16 associations are statistically compelling, the mentioned alleles confer only a small risk for disease and their frequencies in this adult sample are the same as in children, likely limiting their clinical significance. Further epidemiological and functional studies may elicit factors triggering disease in interaction with the risk alleles.Congenital heart disease is the most common birth defect in newborns and the leading cause of death in infancy, affecting nearly 1% of live births. A locus in chromosome 4p16, adjacent to  Expected long-term survival rates have considerably raised with advances in diagnostic and surgical techniques3; survivors require lifelong treatment, posing a growing resource challenge for healthcare systems2. Quantitative genetic research supports a strong genetic contribution to CHD4, and epidemiological findings on survivors have evidenced multiple cardiac and non-cardiac comorbidities that pose novel treatment challenges6.Affecting nearly 1% of live births, congenital heart disease (CHD) is the most common congenital anomaly in newborns and the principal cause of death in infancyde novo mutations7. Building on the current knowledge on CHD genetics, recent large-scale genome-wide association studies (GWAS) have identified some genetic variants increasing the risk for CHD-related phenotypes. In the current GWAS literature, the single most robust association between common genetic variation and CHD risk is a locus in chromosome 4p16, which was initially shown to confer risk for atrial septal defect (ASD)8. This finding was originally reported by Cordell et al.8 in two European-ancestry samples: a discovery cohort of 340 individuals with ostium secundum ASD and 5,159 controls, and a further 417 secundum ASD cases with 2,520 controls. Independent studies in Chinese cohorts have provided additional support to that finding: Zhao et al.9 replicated the results in Han Chinese subjects ; next, a similar outcome was observed in a sample from Southwest China (190 ASD cases and 225 controls)10; and more recently, Pei et al.11 reported analogous findings in a Fujian Chinese population (354 non-syndromic ASD cases and 557 controls).Established molecular genetic contributions to CHD include aneuploidies, copy number variants, point and The six samples mentioned above (two European- and four Chinese-ancestry) thus show consistent evidence of an association between the 4p16 locus and the ASD subtype of CHD –but no clear link to other CHD types–. These observations were based on datasets from either pediatric patients or young adults (median ages up to ~20 years). In this study, the 4p16 susceptibility locus is tested in relation to CHD in older adult survivors: left ventricular outflow tract obstruction (LVOTO) and ASD in a harmonized set of European-ancestry individuals, via the UK Biobank . Ethical approval for the UK Biobank was granted by the NHS National Research Ethics Service (ref: 11/NW/0382).https://biobank.ctsu.ox.ac.uk/).The dataset included in the adult CHD analyses was extracted from the UK Biobank, a longitudinal cohort study of over 500,000 individuals aged 40–69 years. Between 2006 and 2010, participants were enrolled in study centers across England, Scotland and Wales, and extensive baseline data on health behavior, medical history and physical measurements were collected through clinical examination and questionnaires. Similarly, blood, urine and saliva samples were collected from the participants. Ethical approval for the UK Biobank was granted by the North West Multi-Centre Research Ethics Committee and the National Health Service (NHS) National Research Ethics Service (ref: 11/NW/0382). All participants provided written informed consent to participate in the UK Biobank study, and all experiments were performed in agreement with relevant regulations and guidelines. Additional information, including details about the study protocol, is available online .CHD definitions and exclusion criteria were elaborated by leveraging information from medical histories, International Classification of Disease 9 and 10 (ICD-9 and 10) and Office of Population and Censuses Surveys (OPCS-4) diagnosis codes, in addition to information on age at diagnosis and age at surgery. The phenotype classification procedure consisted of four main steps: identification of all possible cases based on diagnostic and procedural codes; exclusion of unconfirmed cases for those who met criteria for non-congenital etiologies of heart disease ; confirmation of case status for those whose age at diagnose suggested high likelihood of congenital etiology, or exclusion otherwise; and exclusion of control group for participants with CHD-related conditions due to lack of confidence to ascertain their status. Briefly, LVOTO patients were identified as possible cases of non-syndromic CHD with LVOTO, and unconfirmed cases were excluded if they met criteria for non-congenital etiologies of heart disease. Then, cases remained unconfirmed when non-CHD etiologies were not found, and were reclassified as cases based on age criteria. Participants were excluded from the control group when there was evidence of conditions associate with CHD. This information, as well as additional details about phenotype definitions are included in Supplementary Methods and Supplementary Fig. R2 between 0.969 and 0.994, as estimated by PLINK 1.9’s–r2 command). The previous observation is in line with the presence of variants within a 33,767-base interval, previous and current association results on any of the four markers may thus represent a single signal. To avoid discussion of redundant analyses, the main marker reported by Cordell et al.8 (rs870142) was used as tag SNP in the meta-analysis moving forward.Demographic data on sex and age distributions across diagnoses is summarized in Supplementary Table 13 indicated that, under an additive model with a prevalence of 1%, alpha of 0.05 and MAF of 23.8%, power to detect an effect size of 1.46  was 88.6% for the UK Biobank’s LVOTO group, and 77.5% for the ASD phenotype. Table n = 167 cases from the larger 223-individual set, ORs between 1.47 and 1.5). Potential survival bias effects were tested by means of logistic regression with the addition of a genotype × age interaction term. As indicated in Fig. Power calculations using CaTSp < 10−4) . Publication bias was not observed in either a standard funnel plot . However, the contour-enhanced funnel plot . Despite this observation, effect sizes observed in the adult ASD survivors from the UK Biobank (OR around 1.3–1.45) were similar to the ones observed in pediatric cohorts . In addition, the 4p16-CHD associations (either LVOTO or ASD) were not modulated by age in this sample  and 332,788 controls; this gives a prevalence estimate of 0.25%, which is much smaller than the classically reported ~0.8–1% for CHD in newbornset al.8), most of them did not include results on LVOTO patient data. A similar diagnostic patient set with left heart malformations was available in Cordell et al.’s8 data, comprising 412 subjects with coarctation of the aorta, hypoplastic left heart syndrome and aortic and mitral valve anomalies, but there was no significant association with the 4p16 risk alleles. This discrepancy between those results and the current analysis in the UK Biobank might be due to clinical and demographic differences: the former patient group were children sampled at a clinical setting, with potentially severe conditions, whereas the UK Biobank includes only adult survivors from the general population. Also, although statistically significant at p < 0.05, the strength of the 4p16-LVOTO association had a smaller effect size and higher p-value than the statistics observed for ASD in the UK Biobank, potentially indicative of weaker effects than those observed for ASD. Having noted those potential limitations, Ellesøe et al.14 recently showed that distinct groups of cardiac malformations co-occur in families with a history of CHD, and concluded that there might be shared susceptibility genes for different CHD subtypes. In line with that, the present results provide preliminary evidence of shared genetic risk for LVOTO and ASD.To the knowledge of authors, this is the first report on the association between the 4p16 locus and the LVOTO phenotype. While some previous studies on the association between the 4p16 locus and ASD also tested for associations with other CHD types , and 53% in their replication cohort (at OR of 1.2). Power was higher in the study by Zhao et al.10 (~69%); however, they reported an unusual MAF of 28.7% in their controls as opposed to the 33% values reported by the other Chinese groups, which may indicate strong differences in ancestry of the participants. The design by Pei et al.11 was well powered to detect the effect they found . Overall, pooled effect size estimates from the meta-analysis should be interpreted with caution, considering the evidence of bias from the contour-enhanced funnel plot. Studies with larger sample sizes are needed to provide better estimates of the underlying effect size.Meta-analysis of ASD from six previously published data and the UK Biobank outcomes had an overall OR of 1.35 for the rs870142 risk alleles in an additive model . Although a simple funnel plot and regression test do not suggest evidence of publication bias . As the number of adult CHD survivors is increasing with the advent of new treatments, the prevalences of different CHD phenotypes will likely change in future studies, limiting between-study comparisons. Moreover, since this is a registry-based dataset and the diagnostic pipeline relied on ICD and OPCS codes, the likelihood of misclassifications and the extent of disease heterogeneity cannot be accurately estimated. Datasets from adult CHD patients recruited and phenotyped directly at clinical settings could help confirm and expand the current findings. Regarding the ASD meta-analysis, attention should be paid to the moderate number of available cohorts (seven independent samples), their modest numbers of ASD cases, within-study power and potential between-study diagnostic differences.Overall, results of this study support an association between the 4p16 locus and CHD, both in pediatric ASD patients and in adult survivors of ASD and LVOTO. Although further research with large and well characterized patient samples is encouraged, only mild effects (OR~1.35) could be expected based on the current evidence. Hence, the clinicopathological significance of these risk alleles does not seem high, and they may only trigger a disease phenotype in the presence of other risk factors. Additional studies are also needed to clarify why the risk alleles are equally frequent in pediatric and adult CHD cases, and to explain why they predispose only to certain CHD subtypes.Supplementary Material


Iron has been shown to promote breast carcinogenesis in animal models through generation of oxidative stress and interaction with estrogen. Heme iron, which is found exclusively in animal-sourced foods, is suggested to have a more detrimental effect. Epidemiological evidence of the association between iron and breast cancer risk remains inconclusive and has not been comprehensively summarized. This systematic review and meta-analysis evaluated associations between both iron intake and body iron status and breast cancer risk.Four electronic databases  were searched up to December 2018 for studies assessing iron intake and/or biomarkers of iron status in relation to breast cancer risk. Using random-effects meta-analyses, pooled relative risks (RRs) and 95% confidence intervals (CIs) were calculated comparing the highest vs. lowest category of each iron measure. Dose-response meta-analyses were also performed to investigate linear and nonlinear associations.Pnonlinearity < 0.05).A total of 27 studies were included in the review, of which 23 were eligible for meta-analysis of one or more iron intake/status measures. Comparing the highest vs. lowest category, heme iron intake was significantly associated with increased breast cancer risk, with a pooled RR of 1.12 (95% CI: 1.04–1.22), whereas no associations were found for dietary , supplemental , or total  iron intake. Associations of iron status indicators with breast cancer risk were generally in the positive direction; however, a significant pooled RR was found only for serum/plasma levels (highest vs. lowest) of iron , but not for ferritin , transferrin saturation , or total iron-binding capacity . In addition, a nonlinear dose-response was observed for heme iron intake and serum iron  iron-rich foods. In light of methodological and research gaps identified, further research is warranted to better elucidate the relationship between iron and breast cancer risk.The online version of this article (10.1186/s12885-019-5642-0) contains supplementary material, which is available to authorized users. Iron is an essential nutrient required for many biological processes in the human body, such as oxygen transport, DNA synthesis, and energy production . HoweverN-nitroso compounds  form) from ferritin storage, further amplifying oxidative stress and inducing DNA damage /progesterone receptor [PR]) tumour subtype were also extracted.Data extraction was performed by one author (VCC) and verified independently by another (EK). For studies with missing information, we referred to related publications  or contacted the corresponding author of the original study for clarification or additional information.The quality of included studies was assessed independently by two authors (VCC and EK) using the Newcastle-Ottawa Scale (NOS) , with anMeta-analyses of the associations between iron intake/status and breast cancer risk were performed separately for each subtype of iron intake or iron status indicator with at least two available studies. When multiple publications reported data on the same iron measure from identical or overlapping study populations, only the publication with the largest sample size or longest duration of follow-up was included in the meta-analysis for the specific iron measure.For each subtype of iron intake or iron status indicator, the pooled relative risk (RR) was used as the summary measure of association and was estimated by combining odds, risk, and hazard ratios reported by individual studies. Odds, risk, and hazard ratios, hereafter all referred to as RRs, were assumed to be equivalent in our analyses given that breast cancer is a relatively rare disease outcome  . If a stIn our main analysis, the pooled RR comparing the highest to the lowest category of each iron intake/status measure was computed. For supplemental iron intake, we examined the dichotomous measure “use vs. no use” instead, as only one study reported RRs across doses of supplemental iron ; for thaTo investigate linear and nonlinear dose-response relations between iron intake/status and breast cancer risk, we further conducted random-effects dose-response meta-analyses using a generalized least-squares method for trend estimation, as proposed by Greenland and Longnecker  and OrsiP < 0.10 considered statistically significant) and the I2 statistic quantifying the proportion of the total variability attributable to heterogeneity [2 values of 25, 50, and 75% roughly indicate low, moderate, and high heterogeneity, respectively [P < 0.10 indicating statistical significance [Heterogeneity between studies was assessed using the Cochran’s Q test . Finally, influence of individual studies was investigated by recalculating the pooled RR and 95% CI each time a single study was omitted from the analysis.Publication bias was evaluated using funnel plots and Begg’s rank-correlation  and EggeP < 0.05 unless otherwise specified.Analyses were performed using Stata/MP, version 14 . Statistical tests were two-sided, with statistical significance evaluated at Our search initially yielded 7589 records. After duplicates were removed, titles and abstracts of 4411 articles were screened, of which 167 full-texts were further assessed for eligibility. Twenty-seven studies, including 17 studies examining iron intake , 52–66 an = 12) [n = 1) [n = 9; including two in Germany and one in each of Denmark, Italy, the United Kingdom, Switzerland, France, Finland, and Sweden) [n = 4; including two in China, one in Taiwan, and one in Japan) [n = 1) [Table n = 12) –70, 73 o [n = 1) , while t Sweden) , 72, 75,n Japan) , 61, 71, [n = 1) .Table 1SOf the 17 studies assessing iron intake, seven were cohort studies , 62–66, With the exception of one cohort study where multiple 24-h dietary recalls were completed during the first 2 years of follow-up , all iroOf the 11 studies examining iron status, nine assessed one or more serum/plasma biomarkers  , 71–75, Results of almost all included studies were reported as RRs across quantiles  of iron intake or status. While age was matched and/or adjusted for in all studies, the level of adjustment of other potential confounders differed across studies Table . Most irDetails of the quality assessment of individual studies are presented in Additional file Figure 2 = 55%, Pheterogeneity = 0.01). With the exception of the one study reporting intake during preschool age  [2 = 56%, Pheterogeneity = 0.02).A meta-analysis combining estimates from 11 studies –62, 65 dctively) , dietaryctively)  was excl2 = 0%, Pheterogeneity = 0.61) and total iron , based on results combined from three studies [Similarly, no associations were found for intakes of supplemental iron . This association persisted after excluding the two studies where animal [2 = 0%, Pheterogeneity = 0.78), or when restricting to studies that used a previously developed laboratory-based database to assess heme iron intake [2 = 0%, Pheterogeneity = 0.94). A meta-analysis was not conducted for non-heme iron intake, as only one study reported its association (assessed as plant-derived iron) with breast cancer risk  [In contrast, heme iron intake showed a significant positive association with breast cancer risk based on six studies , 65, 66,e animal  or red m75–1.29) .Pdifference > 0.10 from meta-regression), with substantial heterogeneity remaining within most subgroups. However, when stratified by study quality, a significant inverse association was observed for low-quality studies , whereas a positive but nonsignificant association was seen for high-quality studies  (Pdifference = 0.03), suggesting study quality may be a contributor to heterogeneity. Furthermore, post-hoc subgroup analyses stratifying results by the highest dietary iron intake category  and method of energy adjustment  did not reveal significant differences . No notable differences were observed for total iron intake, which remained unassociated with breast cancer risk across subgroups. For heme iron intake, all six studies were of high-quality, and significant positive associations remained among cohort studies and studies conducted in North America. In addition, heme iron intake showed a slightly stronger association with premenopausal  than postmenopausal  breast cancer, although statistical significance was not reached in either subgroup.Table Pnonlinearity = 0.41; Fig. Pnonlinearity = 0.46; Fig. Pnonlinearity = 0.03; Fig. Similar to the highest vs. lowest analysis, linear dose-response meta-analyses . The associations were also in the positive direction, but not statistically significant, for ferritin  and TSAT , based on six RRs from five studies [2 = 0%, Pheterogeneity = 0.62). Similarly, serum transferrin was not associated with breast cancer risk according to the only study reporting this measure  [2 = 0%, Pheterogeneity = 0.46; data not shown).Figure  studies , 73–75 a studies , 74, 75, studies  were comTable Pnonlinearity = 0.70)   Fig. a. On the01) Fig. b. SpecifPinteraction = 0.08). In another study where iron levels were measured in benign breast tissue among women with BBD, an elevated breast cancer risk was observed overall  and in postmenopausal women  [Two nested case-control studies assessed iron biomarkers in samples other than serum or plasma , 70. In P = 0.48; Egger’s P = 0.91), supplemental iron , total iron , and heme iron  intake, or for serum/plasma ferritin , iron , or TSAT , whereas publication bias was detected for the combined analysis of TIBC and transferrin . Visual inspection of the funnel plots .No publication bias was detected in the highest vs. lowest meta-analyses for dietary iron  level of iron, but not ferritin, transferrin, TIBC, or TSAT, also showed a statistically significant association with increased breast cancer risk (22%). Furthermore, dose-response meta-analyses indicated a nonlinear threshold effect for heme iron intake and a J-shaped pattern for serum iron in relation to breast cancer risk.This is the first systematic review and meta-analysis specifically assessing breast cancer risk in relation to various measures of iron intake and body iron status. Our review identified many additional studies not included in the previous systematic review/meta-analysis on iron and cancer risk , which oN-nitrosation and lipid peroxidation, and subsequent oxidative damage to cellular biomolecules, have been suggested to contribute to the development of both colorectal and breast cancer [In contrast to our finding of a positive association between heme iron intake and breast cancer risk, the previous meta-analysis reported a lack of association between heme iron intake and breast cancer risk  based on only three studies . The inct cancer , 78. Furt cancer . Surrount cancer  and is at cancer –82 compat cancer .Interestingly, heme iron intake exhibited a nonlinear threshold effect in our dose-response meta-analysis, although absolute intake values should be interpreted with caution given differences in methods used to assess heme iron levels across studies. For example, while literature-based meat-specific percentages  , 14 wereNo overall association was observed between dietary iron intake and breast cancer risk, although results were heterogeneous, with studies reporting positive , 65, invOur meta-analysis provided no evidence of an association between supplemental or total iron intake and breast cancer risk; however, the dose-response curve for total iron intake Fig. b suggestIn contrast to the lack of association observed for total and dietary iron intake, high levels of serum iron were found to be associated with increased breast cancer risk. Similarly, although based only on single studies, iron measured in toenail  and beniThe significant positive association observed for serum iron but not ferritin in our meta-analysis has also been reported by one study examining multiple iron biomarkers within a single population , suggestAlthough stratified analyses by menopausal status generally revealed no significant associations due to limited statistical power, several indicators of iron status  appeared to be more strongly associated with increased postmenopausal breast cancer risk. A possible explanation is age-related dysregulation of iron metabolism and declines in antioxidant defense mechanisms . ConversThis is the first systematic review and meta-analysis specifically assessing the associations between various measures of iron intake, as well as body iron status, and risk of breast cancer. A major strength is the extensive search strategy, allowing us to identify many additional studies not included in the 2014 review on iron and cancer risk , especiaHFE gene underlying hereditary hemochromatosis (iron overload) have been implicated in several cancers, including breast cancer [Several limitations should be considered when interpreting the findings of this review. First, the meta-analysis for some iron measures was based only on a small number of studies, which could have resulted in limited statistical power for the overall or subgroup analyses and the assessments of publication bias, as well as greater influence of single studies. Nevertheless, sensitivity analyses with individual studies omitted one at a time generally led to no notable changes in the pooled estimates, indicating the robustness of our results. Second, our restricted cubic spline dose-response analyses were limited to the range of exposure values derived from individual studies, and the trends observed may be driven by single studies with more extreme exposure values. Thus, a larger number of homogeneous studies across a wide range of exposure values are needed to confirm these results in the future. Third, as an inherent issue in meta-analyses, our analyses combined risk estimates across studies with different designs, populations, settings, statistical adjustments of covariates, etc. , which likely contributed to heterogeneity in our results. Finally, genetic association studies were not considered. For example, mutations in the t cancer . AlthougThis systematic review and meta-analysis suggests that heme iron intake and serum iron levels may be positively associated with breast cancer risk, whereas no associations were found for intakes of dietary, supplemental, or total iron, or serum/plasma levels of ferritin, TIBC, or TSAT. Although the increases in risk were modest, our findings may have public health implications given the widespread consumption of (heme) iron-rich foods. In light of the methodological and research gaps identified in our review , further research is needed to better elucidate the association between iron intake/status and risk of breast cancer.Additional file 1:Electronic database search strategy. (DOCX 17 kb)Additional file 2:NOS coding manuals for study quality assessment. (DOCX 21 kb)Additional file 3:Table S1. Quality of included studies assessed using the NOS. (DOCX 29 kb)Additional file 4:Figures S1 and S2. Linear dose-response analyses of associations between iron intake/status and breast cancer risk. (PDF 298 kb)Additional file 5:Figures S3 and S4. Funnel plots for the evaluation of publication bias. (PDF 307 kb)Additional file 6:Figures S5–S8. Sensitivity analyses investigating the influence of individual studies. (PDF 210 kb)


TOX3 gene plays a key role during the onset of breast cancer, and reproductive factors such as abortion are risk factors for breast cancer. However, there is currently no study describing the relationship between the TOX3 rs3803662 C>T polymorphism and the risk of recurrent miscarriage. Therefore, we investigated whether the TOX3 rs3803662 C>T polymorphism is associated with recurrent miscarriage susceptibility in this case‐control study.Studies have shown that some genetic polymorphisms associated with breast cancer susceptibility may also be associated with abortion. The We recruited 248 recurrent miscarriage patients and 392 healthy controls from the southern Chinese population and performed genotyping using the TaqMan method.TOX3 rs3803662 C>T is associated with recurrent miscarriage .The results showed no evidence that TOX3 rs3803662 C>T polymorphism may not be associated with recurrent miscarriage in the southern Chinese population. A larger multicenter study is needed to confirm the results.According to this study, the  However, there is currently no study on the relationship between the TOX3 rs3803662 C>T polymorphism and the risk of RM. These findings prompted us to evaluate whether the rs3803622 variant of the TOX3 gene is associated with RM susceptibility and whether rs3803622 C>T can be used as a biomedical indicator or a potential risk factor.Located on 16q12.1, the 22.1This study included 248 women diagnosed with RM at the Guangzhou Women and Children's Medical Center between June 2017 and July 2018. Diagnostic criteria were defined as two or more consecutive pregnancy losses before 20 weeks of gestation.2.2All blood samples were supplied by the Clinical Biological Resource Bank of the Guangzhou Women's and Children's Medical Center. Total genomic DNA was extracted from blood using a TIANamp Blood DNA Kit . An ABI Q6 instrument  and TaqMan assays were used to genotype the rs3803622 polymorphism according to the real‐time polymerase chain reaction protocol. The typing probe rs3803622 was purchased from Applied Biosystems (Applied Biosystems TaqMan). The ID number of rs3803622 is ID: C__25968567_10.2.3P values <.05 were considered statistically significant).Hardy‐Weinberg equilibrium (HWE) in the control group was confirmed by the goodness‐of‐fit chi‐square test. Unconditional univariate logistic regression analysis was applied to analyze genotypic and demographic differences between the RM patients and healthy individuals. Furthermore, the association between rs3803622 C>T polymorphism and susceptibility to RM was assessed by the 95% confidence interval (CI) and odds ratio (OR). A stratified analysis of age and number of abortions was performed, and all statistical tests were bilateral. SAS software  was employed for all statistical analyses .The demographic characteristics of the RM patients  and healthy controls are listed in Table 3.2TOX3 rs3803662 C>T polymorphism in RM patients and controls is shown in Table TOX3 rs3803662 C>T genotypes (HWE = 0.636). However, there was no significant association between the TOX3 rs3803662 C>T polymorphism and age‐adjusted RM susceptibility .The genotype distribution of the 3.3TOX3 rs3803662 C>T polymorphism in RM patients and controls  and susceptibility to this condition.Genetic susceptibility may be involved in the onset of recurrent miscarriage and play an important role in the occurrence and development of RM TOX3 (TNRC9) gene is unclear, many studies have found that the rs3803662 polymorphism is associated with multiple diseases. For example, a genome‐wide association study showed that the TOX3 rs3803662 C>T polymorphism increases the risk of breast cancer in the Chinese Han population.TOX3 rs3803662 C>T was significantly associated with the risk of breast cancer. Interestingly, Liao J et allncRNA CCAT2, which is associated with breast cancer susceptibility, is also associated with recurrent miscarriage, and it is possible that similar to lncRNA CCAT2, the TOX3 gene is also associated with recurrent abortion susceptibility. Nonetheless, in our present case‐control study, we did not find any significant association between the rs3803662 polymorphism of the TOX3 gene and susceptibility for recurrent abortion. This may be related to the different role of the rs3803662 C>T polymorphism in different diseases. Larger sample sizes and multicenter cohorts are needed for verification.Although the function of the MALAT1 gene are more protective for women younger than 35 years old and women with 2‐3 miscarriages than rs619586 AA variants26. However, the present study showed that TOX3 rs3803662 C>T was not associated with RM or with the number of prior miscarriages and age. Miscarriage is more prevalent in women older than 40 years,The study found that as the maternal age of pregnant women increases, the risk of miscarriage increases, and maternal age is an important risk factor for spontaneous abortion.TOX3 gene and susceptibility to RM. The role of the TOX3 gene in fetal implantation, fetal growth maintenance, or gestational immune balance was not investigated. Finally, the sample size in our current study is relatively small. A larger sample size is needed in future studies to confirm the role of rs3803622 C>T in the TOX3 gene in susceptibility to recurrent abortion.The current research has some limitations. First, the study was limited to the southern Chinese population and did not assess cases and controls among other ethnic groups. Given the large differences in the incidence of miscarriage in different races, the distribution and function of rs3803662 C>T in different populations need further study. Second, the analysis of miscarriages included age and the number of abortions, but due to lack of information, other factors such as family history were not considered in the stratified analysis, and such factors may affect the outcome of this study. Third, we only studied the relationship between rs3803622 C>T polymorphism in the TOX3 rs3803662 C>T polymorphism and miscarriages and other parameters, such as the number of abortions and the age at which miscarriages occurred, was not significant. A larger multicenter study is needed to confirm this conclusion.In summary, we conducted a case‐control study with a small sample size of 248 patients and 392 healthy controls from the southern Chinese population. The association between The authors report no conflicts of interest.All authors contributed significantly to this work. DC and WDH devised the research plan. The data were analyzed by YFX and YFY. DC wrote the study, and QSL, ZLL, and HZZ were responsible for performing the experiments. LP and LYF designed the experimental methods, and XQG modified and polished the study. All authors support the publication of the manuscript.


We would like to thank T‐H Kim and JW Kang for carefully reading our paper. The authors state in the title of their letter that there could not be a strong association between acupuncture and HCV infection, a statement that should not be made neither based on our systematic review (SR), where we found a significant association  based on very‐low‐quality evidence, nor on the additional data they provide. We would like to explain this in more detail and further comment on the methodological ‘issues’ that were raised by the authors.Firstly, the authors refer to one single study Secondly, the authors state that we omitted two studies from the meta‐analysis. However, a lack of data was present in 1 study since no information on the controls was reported Lastly, the authors asked to consider risk differences (RD) as effect measure to avoid problems with zero‐event studies when calculating the OR. The resulting forest plot  can be found with this letter, showing that the significant OR now changed into a non‐significant RD of 0.06, 95%CI . However, according to the Cochrane manual, RD methods yield too wide confidence intervals when events are rare, which make them unsuitable for meta‐analysis of rare events In summary, the available evidence does not support the conclusion that there could not be a strong association between acupuncture and transfusion‐transmissible infections. In order to gain more insight in this association, better quality studies are necessary.