Several observational studies on the association between Cd exposure and risk of prostate cancer have yielded inconsistent results. To address this issue, we conducted a meta-analysis to evaluate the correlation between Cd exposure and risk of prostate cancer.Relevant studies in PubMed and Embase databases were retrieved until October 2015. We compared the highest and lowest meta-analyses to quantitatively evaluate the relationship between Cd exposure and risk of prostate cancer. Summary estimates were obtained using a random-effects model.In the general population, high Cd exposure was not associated with increased prostate cancer , whereas the combined standardized mortality ratio of the association between Cd exposure and risk of prostate cancer was 1.66 (95% CI 1.10–2.50) in populations exposed to occupational Cd. In addition, high D-Cd intake  and U-Cd concentration  among the general population was not related to the increased risk of prostate cancer. In the dose analysis, the summary relative risk was 1.07 (95% CI 0.73–1.57) for each 0.5 μg/g creatinine increase in U-Cd and 1.02 (95% CI 0.99–1.06) for each 10 μg/day increase of dietary Cd intake. However, compared with nonoccupational exposure, high occupational Cd exposure may be associated with the increased risk of prostate cancer.This meta-analysis suggests high Cd exposure as a risk factor for prostate cancer in occupational rather than nonoccupational populations. However, these results should be carefully interpreted because of the significant heterogeneity among studies. Additional large-scale and high-quality prospective studies are needed to confirm the association between Cd exposure and risk of prostate cancer. In 2015, up to 220,800 men were diagnosed with prostate cancer, and 27,540 men will die of it in the United States.2 In recent decades, a rapid increase in prostate cancer incidence has been observed in fast-developing countries, in which lifestyles have significantly changed. The etiology of prostate cancer comprises multiple factors. Some causative risk factors for prostate cancer have been implicated, including obesity, androgen, and exposure to selenium, lycopene, vitamins D and E, dietary fat, and Cd.7Prostate cancer is one of the most common malignancies in developed countries and is the second most common cancer in men, following lung cancer, worldwide.8 Besides being a carcinogen,9 Cd exposure is associated with osteoporosis and bone fracture,11 type 2 diabetes,12 kidney disease,13 and cardiovascular disease.15 Cd is currently one of the most extensive occupational and environmental pollutants. Occupational Cd exposure is used in various industries, such as Cd-emitting industries and metal mines. Occupational Cd exposure occurs when dust and fumes are inhaled. In particular, major sources of natural and anthropogenic Cd in the general population include cigarette smoking and diet choices: tobacco, grains, potatoes, and vegetables taking up Cd from the soil.16 Several epidemiologic studies investigating the association between Cd exposure and susceptibility to prostate cancer have yielded inconsistent findings. Some studies have demonstrated a significant correlation22 or little association25 between Cd exposure and risk of prostate cancer, but others failed to show any significant association.39 Therefore, we systematically performed a meta-analysis by combining all available data from observational studies to evaluate the association between Cd exposure and risk of fracture. Our meta-analysis followed the Preferred Reporting Items for Systematic Reviews.40Cd is a minor metal found naturally in the earth's crust and has been widely distributed in the environment as a result of anthropogenic activity. Cd presents an elimination half-life of 10 to 30 years and may exert a wide range of negative effects on human health.40This article presents a systematic review and meta-analysis of previously published studies; therefore, ethical approval and written informed consent from patients are not required. This research was conducted in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analysis statement.We searched the PubMed and Embase databases until October 2015 to identify relevant studies that evaluated the association between Cd exposure and prostate cancer risk. We used the following search terminologies: “prostate carcinoma” OR “prostatic cancer” OR “prostate cancer” OR “prostatic carcinoma” combined with “Cadmium.” The search was limited to human subjects. Moreover, we manually searched the reference lists of previous reviews and related article references to identify other potentially eligible studies.The inclusion criteria were as follows: (1) Cd was the heavy metal used for exposure to humans; (2) the outcome of interest was prostate cancer incidence, prevalence, and mortality; (3) the report was a cohort, case-control study; and (4) the relative risk (RR), odds ratio (OR), hazard risk, or standardized mortality ratio (SMR) with corresponding 95% confidence interval (CI) was reported or calculated from available data. If the study was reported more than once, we included the study with the most comprehensive data.Two authors (JKS and DBY) separately extracted data from selected studies, and discrepancies were resolved through discussion and consensus. For each study, we extracted the first author's name, year of publication, study design, country, total number of cases and subjects, sex, exposure type of Cd, and adjusted variables. When more than 1 adjusted OR was reported, the OR with the most fully adjusted model was selected. For dose analysis, the number of cases and participants of person-years for each category of Cd exposure must also be provided .41 The checklist contained 9 items for case-control studies and cohort studies, with every item accounting for 1 point. We considered high-quality studies as those with a score of >5.We evaluated the methodological quality of the included studies by using the Newcastle–Ottawa scale (NOS).23 and region.38 We combined these estimates by using a random-effects model and used the pooled estimates for the meta-analysis. The OR in 1 study34 was not extracted; thus, we computed the crude risk estimates and their corresponding CI. A random-effects model of the DerSimonian and Laird method was used to calculate the summary risk estimates, irrespective of heterogeneity, which incorporated both within-study and between-study variability.42 Subgroup analysis was stratified by geographic region, study design, quality of NOS scale, type of outcome, and type of exposure. We conducted sensitivity analyses by omitting 1 study in each turn to investigate whether the results were attributed to 1 large study or a study with extreme results. Furthermore, we explored the heterogeneity of the different variables mentioned above through a single-variable meta-regression analysis. We conducted a 2-stage random-effects dose-response meta-analysis by using the method proposed by Greenland and Longnecker.43 This method required that the distribution of cases, person-years, non-cases, and risk estimates within the variance are known for at least 3 quantitative exposure categories. We assigned the median values or middle point of Cd exposure for each category to the corresponding RR. If the highest category of the studies was open-ended, we assumed the range to be the same as the adjacent interval. First, we estimated a restricted cubic spline model via generalized least-square regression with 4 knots at 5%, 35%, 65%, and 95% distribution. Second, we pooled the study-specific risk estimates by using the restricted maximum likelihood method in a random-effects meta-analysis. Nonlinear relation was estimated by testing the null hypothesis, which indicated that the coefficient of the second spline is equal to 0.Differences were expressed as OR with 95% CI for nonoccupational exposure studies and SMR with 95% CI for occupational exposure studies. Prostate cancer caused by Cd was considered a rare event, and the RR in the cohort study was considered as approximations of OR. Three studies reported stratified risk estimates by ageP < 0.1 indicating a significant publication.44 If asymmetry evidence was detected, the trim-and-fill method was employed to correct the publication bias.45 All statistical analyses were conducted using Stata version 13.1 .We also evaluated the potential publication bias by using funnel plot and Egger tests, with a priori 49 4 articles presented no data useful for the meta-analysis,53 2 articles were excluded because the Cd contents in prostate tissues were measured,55 2 articles were reviews,56 and 1 article reported an association between Cd and prostate-specific antigen.57 Finally, 22 studies that met the meta-analysis criteria were included.Figure 39 7 in the United States,34 and 2 in Asia.28 Thirteen studies reported findings for prostate cancer incidence,39 whereas the remaining 9 studies reported results for prostate cancer mortality.38 We included a total of 200 prostate cancer deaths, 6653 prostate cancer cases, and 137,998 participants in the meta-analysis. Eight studies were designed to evaluate OR,36 3 evaluated RR,39 3 evaluated hazard risk,28 and 8 evaluated SMR38. Three articles used urinary Cd (U-Cd) as biomarker for long-term exposure to Cd,27 4 articles evaluated Cd levels by estimating the dietary Cd (D-Cd) by using food frequency questionnaires,39 and 2 studies examined Cd in toenails33. Twelve studies reported an association between occupational Cd exposure and prostate cancer risk,38 whereas 10 studies used nonoccupational populations.39 Most of the studies were controlled for some conventional risk factors, including age (n = 6) and smoking (n = 6). Some studies were also controlled for body mass index (n = 3) and alcohol consumption (n = 2), but few studies were adjusted for beef intake, dairy product consumption (n = 1), and intake of vegetable and fruit (n = 1). None of the studies were adjusted for other heavy metals, trace elements of organic pollutants, and intake of grains.Tables NOS was used to assess the quality of studies included in the meta-analysis Table . The medPfor.heterogeneity = 0.002; I2 = 69.9%). In subgroup analyses for exposure type, we restricted each analysis to 7 occupational exposure studies, resulting in a summary SMR of prostate cancer of 1.61 (95% CI 1.04–2.48). Only 1 study was conducted in the United States,31 and the 7 other studies were conducted in Europe. When we stratified the analysis by geographic region, the pooled SMR was 1.63 (95% CI 1.05–2.52) for studies conducted in Europe. Compared with a low NOS score , the association was higher among studies with high NOS score    to 1.87 (95% CI 1.23–2.83) with significant heterogeneity  . Egger test (P = 0.241) and funnel plot . In subgroup analyses for study design, we restricted each analysis to 9 case-control studies and 5 cohort studies; the summary ORs of prostate cancer for the highest category of Cd exposure versus the lowest category were 1.31 (95% CI 0.60–2.87) and 1.06 (95% CI 0.99–1.09), respectively. Five studies were conducted in the United States, 6 in Europe, and 2 in Asia. When we stratified the analysis by geographic region, the combined OR was 1.10 (95% CI 0.66–1.84) in studies conducted in the United States, 1.57 (95% CI 0.86–2.84) in studies conducted in Europe, and 0.74 (95% CI 0.34–1.60) in studies conducted in Asia. When stratified by type of Cd exposure, the combined OR of prostate cancer was 1.05 (95% CI 0.97–1.15) for D-Cd, 1.18 (95% CI 0.28–2.34) for U-Cd, and 1.87 (95% CI 0.29–12.06) for toenail Cd. Five studies reported an association between occupational Cd exposure and prostate cancer risk; however, the association was not significant in the occupational exposure population . When stratified by type of outcome, the combined OR was 1.23 (95% CI 0.78–1.95) for prostate cancer incidence and 1.29 (95% CI 0.51–3.27) for prostate cancer mortality. Compared with studies that presented low NOS scores , the association was higher among studies with high NOS scores   with moderate heterogeneity . Sensitivity analysis showed that the overall pooled estimate was not altered substantially, with the exclusion of 1 study. The overall combined OR after sequential exclusion of 1 study at a time ranged from 1.03 (95% CI 0.88–1.21) with significant heterogeneity   to 1.29  with significant heterogeneity  . No evidence of publication bias resulted from Egger test (P = 0.881) and near-symmetric funnel plot . The summary RR per 0.5 μg/g creatinine increment was 1.07 (95% CI 0.73–1.57) with no evidence of heterogeneity . Four studies were included in the dose-response analysis of D-Cd intake and prostate cancer risk. We found no significant departure from a simple linear-response association between Cd exposure and prostate cancer (Pfor nonlinearity test = 0.64). The estimated RR of prostate cancer risk was 1.02 (95% CI 0.99–1.06) for 10 μg/d increase of D-Cd, with little evidence of heterogeneity  . Most nonoccupational Cd exposure studies,39 though not all,20 showed no significant relation between Cd exposure and prostate cancer risk. Verougstraete et al67 conducted a systematic review on Cd-exposed workers and found that Cd exposure was not associated with increased risk of prostate cancer. In another recent meta-analysis based on 8 previous studies, in which D-Cd intake showed no statistically significant association with cancer risk except in stratified analysis by geographic region, a positive association between D-Cd intake and cancer risk was observed.56 However, few studies were included in the meta-analysis, which limited the possibility of drawing robust conclusions, especially in the subgroup analysis. Compared with the 2 previous meta-analyses, the current research presented a more extensive systematic review, which included a large number of studies with more than 6828 cases, 123 deaths, and almost 171,972 participants. Thus, we obtained adequate statistical data to clarify the relation between Cd exposure and risk of prostate cancer. In our meta-analysis, we found a positive association between high Cd exposure and risk of prostate cancer for occupational exposure, but not for nonoccupational exposure. These findings can potentially result in higher Cd exposure levels prevailing in these studies.Cd is a nonessential metal widely distributed in the environment by industrial and agricultural activities.No substantial changes were observed in most subgroup analyses because the Cd concentrations in the blood and urine are the most common biomarkers for Cd exposure. U-Cd mainly reflects Cd accumulation in the kidney, as determined by lifelong exposure, whereas D-Cd demonstrates a combination of both current and historical exposure. Results from subgroup analyses stratified by type of Cd exposure showed that both U-Cd and D-Cd were not associated with increased prostate cancer risk. Smoking is a primary source of exposure in the general population and is known to damage health through direct and indirect effects. Thus, we also performed subgroup analyses among studies controlled for smoking status to minimize possible non–Cd-mediated negative effects of tobacco smoking on prostate cancer risk. Six publications were adjusted for smoking status, and the results showed that Cd exposure was not associated with increased prostate cancer risk .69 and numerous experimental studies in vivo and in vitro have indicated that Cd can act as a prostate carcinogen in rats.70 Cd has been recognized as a human carcinogen by the International Agency for Research on Cancer on the basis of mechanistic and epidemiologic evidence from high-exposure occupational settings.9 Several mechanisms are responsible for the carcinogenesis of Cd exposure, including induction of oxidative stress,71 suppression of DNA repair,72 alterations of DNA methylation,73 inhibition of apoptosis proto-oncogene activation,74 tumor suppressor gene inactivation, and cell adhesion disruption.75 In addition, Cd can exert estrogenic activities that play a role in the development of prostate cancer. Cd is suggested to exert estrogenic properties, and direct receptor-mediated effects of estrogen on prostate are plausible.76 Experimental evidence showed that excessive exposure to estrogens can cause prostate cancer. In human prostate epithelial cells, Cd exhibits estrogenic activity, including proliferation of prostate cells and activation of the estrogen receptor-α.77 Substantial evidence showed a positive relation between Cd exposure and risk of endometrial78 and breast cancers.79 Therefore, high Cd exposure can potentially increase the risk of prostate cancer.Many studies have demonstrated that the prostate is a target organ for the deposition of Cd,The present meta-analysis exhibited several strengths. The first research highlight of this meta-analysis is its large sample size. The large number of total cases provided high statistical power to quantitatively evaluate the association between Cd exposure and prostate cancer risk. Second, publication bias is a potential concern in any meta-analysis because small studies with null results do not get published. However, in our meta-analysis, we found little evidence of publication bias.22 that reported a significant positive association and yielded a low NOS. Results from subgroup analyses indicated that geographic region, study design, quality of NOS, type of outcome, and type of exposure are potential sources of heterogeneity. Nevertheless, we used meta-regression and sensitivity analysis to explore the potential causes of between-study heterogeneity. Our meta-regression analysis did not find covariates of publication year, study design, geographic region, NOS, type of outcome, and type of Cd exposure as sources of heterogeneity. Finally, although we selected the highest multivariable-adjusted effect estimates in our meta-analysis, we cannot exclude the possibility that the observed increase in association between Cd exposure and prostate cancer risk among occupational populations can be ascribed to unmeasured or residual confounding factors. The unstable results were observed in occupational and environmental populations, indicating that more relevant articles are needed to further explore this association.Nevertheless, some limitations should be considered in the present meta-analysis. First, observational studies, even if prospective, cannot prove causality. We cannot exclude the possibility that the observed positive relationship between Cd exposure and prostate cancer risk is attributed to confounding factors. Majority of the studies were adjusted for potential confounding factors, but not all potential confounders were adjusted in every study. For instance, a potential confounder such as cigarette smoking is not only a source of Cd but also contains other substances with adverse health effects. In analyses stratified by adjusting the smoking status, similar results were obtained. Most studies were adjusted for some conventional risk factors, including age and smoking status, and some studies were controlled for body mass index and alcohol consumption. However, few studies were adjusted for other dietary variables or nutrients, whereas none of the included studies were controlled for other heavy metals, trace elements, or organic pollutants. Second, an accurate assessment of Cd exposure remains a challenge. Most studies used questionnaires to assess Cd exposure, whereas some research used interviews, company records, and self-reports to evaluate Cd concentration. However, increasing errors in measurements become inevitable. The imprecise measurement of Cd concentration might have attenuated the true associations. Third, the definition of Cd exposure varied across studies. The Cd exposure types differed according to geographical locations, as urine Cd concentration (in μg/g creatinine) ranged from approximately 0.39 to 1.46 in the US and European population. In Asian studies, the mean urine Cd concentration (in μg/g creatinine) ranged from 0.94 to 1.4. The Cd intake from food generally varies between 9 and 25 μg/d in the US and in Europe. In Asian studies, the mean D-Cd intake (in μg/day) ranged from 19.7 to 35.4. These factors can affect our results. However, our subgroup analyses showed that the associations between Cd exposure and prostate cancer risk did not differ significantly in terms of study location. Third, potential sources of between-study heterogeneity, which is common in meta-analyses, should be explored. In sensitivity analyses, the observed heterogeneity was explained by an articleIn summary, this meta-analysis suggests high Cd exposure as a potential risk factor for prostate cancer in occupational populations but not in nonoccupational populations. However, these results should be carefully interpreted because of the significant heterogeneity among studies. Additional large-scale and high-quality prospective studies are needed to confirm the association between Cd exposure and risk of prostate cancer.


The relationship between hypertension and hemifacial spasm (HFS) has been debated. Microvascular decompression surgery is effective in some HFS patients with uncontrolled hypertension. To address current gaps in knowledge, we conducted a meta-analysis of case-control studies that have examined the prevalence of hypertension in HFS patients compared to non-HFS controls. We also evaluated the implications and limitations of the pooled studies. We identified 62 studies from PubMed, The Cochrane Library, Web of Science and Scholar.google.com and six studies that fit our inclusion criteria were included. A random-effects model was used to derive the pooled estimate of the Odds Ratio. The data was plotted on a Forest plot. A pooled analysis involving 51585 subjects, 549 cases, 720 neurological controls and 50316 controls from the general population, showed that HFS patients had a higher chance of developing hypertension , p-value  <0.001). The prevalence of hypertension was higher in HFS patients as compared to non-HFS patients. This meta-analysis highlights a positive correlation between hypertension and HFS. Blood pressure should be closely monitored during the follow-up of HFS patients. Preliminary links between ventrolateral medullary (VLM) compression and HFS should be further evaluated in future studies. Hemi-facial spasm (HFS) is characterized by intermittent, involuntary contraction of facial muscles, usually unilaterally, though it can affect both sidesHypertension is a relatively common ailment among the general population. Based on results from the National Health Survey in SingaporeHypertension has long been suspected as having a cause and effect relationship with HFS, and studies have been conducted to look into this relationship but their findings have been debated. A meta-analysis of properly conducted case control studies will help to address this relationship. Hence, we did a meta-analysis of the case-control studies that have investigated the relationship between hypertension and HFS and evaluated the challenges and limitations of these studies.et al.A total of 549 HFS cases and 720 neurological controls contributed to the meta-analysis. Sandell et al., all controls were age- and gender-matched and there were no significant statistical differences in demographics. The study by Sandell et al. did, however, account for age and gender when comparing the data. Participants with secondary causes of hemifacial spasm and hypertension were also excluded from the study to reduce confounding effect. Notably, the studies by Tan et al.et al.The mean sample size per study for cases was 91.5 while that for neurological controls was 144. The overall gender percentages for males were 33.2% and 32.1% for cases and controls respectively. In cases, the overall age range was from 20 to 83 years old, while in controls, the age range was from 20 to over 90 years old. Within each study, apart from the study by Sandell I2-statistic, pooled estimate of Odds Ratio was derived using a random effects model to account for heterogeneity. The pooled result by means of a random effects model showed that patients had a high chance of developing hypertension , p-value  <0.001). This is presented on the Forest Plot  this was not found to be statistically significant, since the 95% confidence interval straddles 1. However, as the sample size is small, we did not sub-analyse based on ethnicity.All the individual studies produced an Odds Ratio of more than 1.0, for the prevalence of hypertension in cases versus controls. In the Asian studies .The heterogeneity of the study populations may also suggest that there is no “one size fit all” intervention we can apply, and each population would still need to streamline the interventions to best fit their target populationThe results of the meta-analysis showed that there is indeed a higher prevalence of hypertension among HFS patients as compared to non-HFS controls. These findings bring about clinical implications.In patients presenting with HFS without hypertension, we need to have a high index of suspicion that hypertension may be present at diagnosis or develop in the course of the illness. Therefore, regular blood pressure monitoring should be performed during follow-up visits for early detection and intervention of hypertension should it developet al. found that an increased duration of arterial hypertension was a significant risk factor for HFS, suggesting a possible etiologic linket al.et al.In addition, in patients with hypertension but no HFS, we also need to consider that HFS may develop. Although hypertension is not a clear risk factor for HFS, a study by Rudzinska Since HFS is a very uncommon disease compared to hypertension, the risk of developing HFS in a hypertensive patient is still low, it may seem overly cautious to warn every patient with hypertension about HFS. However, it may be good to educate the general practitioner on this link so they can keep a watch out for the patients under their care. Future studies should address if controlling hypertension can help in managing HFS.et al. demonstrated a positive correlation between hypertension and compression of the left-sided medulla oblongata, suggesting a possible etiologic link between VLM compression and hypertension13The link between HFS and hypertension brings ventrolateral medullary (VLM) compression into question. A meta-analysis done by Boogaarts Current case-control studies in HFS and hypertension have limitations. Their cross-sectional design only allowed the prevalence to be measured from a single point in time. A prospective cohort study with less bias and confounding factors may provide additional clues on the cause and effect associationIn conclusion, this meta-analysis highlights a positive correlation between hypertension and HFS. Blood pressure should be closely monitored during the follow-up of HFS patients. Preliminary links between VLM compression and HFS should be further evaluation in future studies.Case-Control StudyData publicly available for analysisSample size of at least 60 HFS casesPublished from Year 2000 onwardsPapers were sourced from The Cochrane Library, PubMed, Web of Science as well as Scholar.google.com. We identified 62 papers and eventually shortlisted 7 of these papers based on the following inclusion criteria:We chose to use case-control studies where a clear comparison is made between the prevalence of hypertension in HFS cases and that in the control group. The data could then be used to generate the values to be used in the meta-analysis. While publication bias may arise from only using publicly available data, we ensured that we screened through all the abstracts, even those from journals which were not publicly available, and found that none of these studies were relevant to our topic. 2 papers11Meta-analysis helps to increase the statistical power of the data, thus increasing the accuracy of the results drawn as compared to each individual study. This is primarily achieved through increasing the sample size, by combining the data from multiple papers studying the same thing and calculating the weighted average of the results. In addition, by compiling data from these papers and comparing the results on a single Forest plot, convergences and divergences between the results from different papers are represented clearly in a visual format and further analysis can be madeQ-statistic was used to investigate the degree of heterogeneity between studies. The overall inconsistency index, I2-statistic, was evaluated to describe the proportion of total variation caused by heterogeneity. I2-statistic of less than 30% of the variability in point estimate was considered as mild heterogeneity, more than 50% was notable heterogeneity, whereas in between was considered moderate heterogeneity.Only case-control studies were included and the Odds Ratio reflected on a Forest plot. How to cite this article: Leong, J.-L. et al. Revisiting the link between hypertension and hemifacial spasm. Sci. Rep.6, 21082; doi: 10.1038/srep21082 (2016).


Epidemiological studies have reported conflicting results regarding maternal parity and the risk of congenital heart defects (CHDs). However, a meta-analysis of the association between maternal parity and CHDs in offspring has not been conducted.We searched MEDLINE and EMBASE for articles catalogued between their inception and March 8, 2014; we identified relevant published studies that assessed the association between maternal parity and CHD risk. Two authors independently assessed the eligibility of the retrieved articles and extracted data from them. Study-specific relative risk estimates were pooled by random-effects or fixed-effects models. From the 11272 references, a total of 16 case-control studies and 3 cohort studies were enrolled in this meta-analysis.Q = 32.34; P = 0.006; 2I = 53.6%). Furthermore, we observed a significant association between the highest versus lowest parity number, with an overall RR = 1.20 . A dose–response analysis also indicated a positive effect of maternal parity on CHD risk, and the overall increase in relative risk per one live birth was 1.06 ; Q = 68.09; P<0.001; 2I = 80.9%). We conducted stratified and meta-regression analyses to identify the origin of the heterogeneity among studies. A Galbraith plot was created to graphically assess the sources of heterogeneity.The overall relative risk of CHD in parous versus nulliparous women was 1.01 (95% CI, 0.97–1.06; In summary, this meta-analysis provided a robust estimate of the positive association between maternal parity and risk of CHD. Congenital heart defects (CHD) are the most common human birth defects and the leading cause of perinatal mortality, with an incidence of approximately 4 to 50 per 1000 live birth or even higher Maternal phenylketonuria, diabetes mellitus, maternal teratogen exposure, and maternal therapeutic drug exposure during pregnancy may increase the risk of congenital malformations in offspring To date, an increasing number of studies has focused on the association between maternal parity and CHDs; however, the results have been ambiguous, possibly because of inadequate sample sizes. Therefore, we conducted a dose-response meta-analysis to quantitatively assess the effects of maternal parity on CHDs.To identify relevant epidemiological studies, two independent researchers (Feng and Yu) conducted a computerized literature search in MEDLINE and EMBASE to retrieve articles that were catalogued between the databases' inception and March 8, 2014.The search terms for the exposure were: ‘Parity’, ‘Pregnancy’, ‘Live Birth’, ‘Reproduction’, ‘Reproductive’ and ‘Reproductive Factors’ and the search terms for the outcome were: ‘Congenital Heart Defect’, ‘Heart Abnormality’, ‘Malformation Of Heart’ and ‘CHD’. In addition, we conducted a search for a broad range of environmental teratogens and CHDs and examined the relevant references and review articles; in this way, we could identify information from other related studies. We followed standards of quality for conducting and reporting meta-analyses We selected articles that (1) were original epidemiologic studies , (2) examined the association between maternal parity and CHDs overall or any one of the CHD subtypes in infants, (3) were published in the English language, (4) reported RRs  and associated 95% confidence intervals (CIs) or standard errors or provided the data necessary to recalculate these factors, and (5) defined CHDs or one of the CHD subtypes as an outcome. Articles that reported results from more than one population were considered to be separate studies. When multiple articles from the same study were provided, we used the article with the most applicable information and the largest number of cases. We excluded non-peer-reviewed articles, experimental animal studies, ecological assessments, correlation studies and mechanistic studies.Data extraction was carried out separately by two reviewers (Feng and Yu) working independently. When differences of opinion arose, they were resolved by a discussion between the two reviewers or by the involvement of a third reviewer (Chen) for adjudication. Parity was defined as the number of live births before the index delivery To assess the study quality, we used a 9-star system on the basis of the Newcastle-Ottawa Scale We used study-specific relative risks as a summary statistic of the association between maternal parity and CHD risk. To simplify the procedure, a RR was used to represent all reported study-specific results from cohort studies and an OR to represent results from case-control studies. If a study did not use the lowest parity number as the reference category, the effective count method proposed by Hamling and colleagues For the dose–response analysis, which considers parity as a continuous variable, the method proposed by Greenland and colleagues Q and 2I statistics were used to test for heterogeneity among studies P<0.05 or 2I≧56%), a random-effects model was used, which provided a more appropriate summary estimate for heterogeneous study-specific estimates. If the study revealed no evidence of heterogeneity, the fixed-effects analysis was used, an inverse variance weighting was applied to calculate summary RR estimates Cochran's P value less than 0.05 from the meta-regression was considered representative of a significant difference between subgroups. Finally, we conducted sensitivity analyses to explore whether a specific study strongly influenced the results by excluding one study at a time.We conducted subgroup analyses based on study design , geographical region , number of cases , publication period , maternal age , primary interest , and study quality . We evaluated heterogeneity between subgroups by meta-regression. A P value of <0.05 for the two above-mentioned tests. All statistical analyses were performed with STATA .Publication bias was assessed via visual inspection of a funnel plot with asymmetry using both Egger's linear regression The search strategy generated 11272 citations; from these, 17 were used in the final analysis, representing 43880 incident cases . All of Q = 32.34; P = 0.006; 2I = 53.6%; P = 0.295) or visual inspection of the funnel plot (data not shown). In a sensitivity analysis, we sequentially excluded one study at a time and reanalyzed the data. The 16 study-specific relative risks for the parous versus nulliparous women ranged from a low of 1.01  after omission of the study by Padula and colleagues Q = 31.44; P = 0.018; 2I = 45.9%) after omission of the study by Luo and colleagues P for heterogeneity = 0.12).A total of 14 case–control studies and 2 cohort studies examined the association between parity as a binary variable and CHD risk. The overall relative risk of CHD for parous versus nulliparous women was 1.01 , with moderate heterogeneity . The 13 study-specific relative risks when considering the parity number ranged from a low of 1.17  after omission of the study by Vereczkey and colleagues Q = 59.74; P = 0.000; 2I = 79.9%) after omission of the study by Batra and colleagues A total of 11 case–control studies and 3 cohort studies examined the association between high and low parity and CHD risk. The estimate of the relative risk of CHD for the highest versus lowest parity categories was 1.20 . Statistically significant heterogeneity was detected  or Begg test (P = 0.101), and no asymmetry was observed in the funnel plots. The 13 study-specific relative risks of parity ranged from a low of 1.05  after omission of the study by Vereczkey and colleagues Q = 51.12; P = 0.000; 2I = 76.5%) after omission of the study by Cedergren and colleagues A total of 11 case–control studies and 3 cohort studies were included in the dose-response analysis. The estimate of relative risk per live birth was 1.06 , and there was statistically significant heterogeneity . We further created a Galbraith plot to graphically assess the sources of heterogeneity ; however, the corresponding pooled RRs were not materially altered in any comparisons .We conducted stratified and meta-regression analyses to identify the origin of the heterogeneity among studies. In subgroup analyses of parity as a binary variable and CHD risk, there was no indication of significant heterogeneity between subgroups according to meta-regression analyses Table 1.ogeneity , S2. A tTo the best of our knowledge, this is the first quantitative meta-analysis evaluating the association between maternal parity and the risk of congenital heart defects. Overall, the findings of our meta-analysis suggested that maternal parity  was significantly associated with CHD risk. Meanwhile, in the dose-response meta-analysis, we found that the risk of CHD increased by 6% per live birth. However, there was no evidence that verified the association between parous versus nulliparous women  and the risk of CHDs. Additionally, the results were consistent across most of the subgroup analyses  and 2.Although the specific biological mechanism underlying maternal parity and the risk of CHDs remains unclear, some relevant evidence has been published. Nutrient depletion was more likely to occur among mothers who had given birth to live fetuses than those who had never delivered. Folic acid is one of the most important vitamins, and the association between folic acid and birth defects has been widely studied. It has been confirmed that lack of it would cause severe congenital malformation When stratified by geographic region, a significant increase in CHD risk in North America and Europe was found to be associated with increases in parity number, and similar results were found in a dose-response meta-analysis. However, the pooled RRs for North America and Europe differed when considering parity as a binary variable. Considering the fact that only one study from Asia was included, the influence of parity in this region needs further research. In the subgroup analysis to assess study quality, we observed statistically significant results in high quality studies that included analyses of both parity number and dose-response, while no significant association was found among low quality studies. For the subgroup analysis of study design, the pooled RR from case-control studies was different from cohort studies in the analysis of dose-response. Selection and information biases might account for the observed difference. Furthermore, compared to the cohort studies, the case-control studies had a lower median quality score (7 versus 8), which may have an influence on the results.P for heterogeneity = 0.12 in maternal ever parity; P for heterogeneity = 0.106 and P for heterogeneity = 0.157 in maternal parity number). So we consider that maternal age may have no significant confounding effect on association between maternal parity and CHD in offspring. However, because of the limiting number of included studies, more studies are needed to validate our results.Some limitations of our study must be taken into account. First, a total of 14 case-control studies and 3 cohort studies were recruited into our meta-analysis, and we extracted our raw data primarily from case-control studies, which are susceptible to selection and information biases. Additionally, our meta-analysis was limited to studies published in English; the results may therefore have been affected by the lack of data from studies performed in other languages. Thus, general conclusions must be considered carefully and cannot be regarded as the final word on the matter. Second because we lacked a large data set, we did not conduct a subgroup analysis of CHD subtypes; however, different CHD subtypes have different etiologies. Maternal parity may be not associated with all subtypes of CHD. Therefore, further research, including more high quality studies, is needed. Thirdly, although no evidence of publication bias was found, heterogeneity exists among the studies included in these analyses of both parity number and dose response; this heterogeneity may affect the interpretation of the overall results. In this study, we conducted sensitivity analyses to explore the sources of heterogeneity by deleting one study at a time from the pooled analysis. However, heterogeneity still could not be fully removed. Moreover, geographical region, sample size, CHD subtypes and other risk factors may result in heterogeneity. Therefore, we performed meta-regression and subgroup analyses to further investigate the sources of heterogeneity. In the dose-response analysis, we found that the heterogeneity stemmed partly from the number of cases. In contrast, no cause was found for the heterogeneity in the parity number meta-analysis. Furthermore, we created a Galbraith plot to assess the heterogeneity and to identify potentially outlying studies. A total of 6 were identified as the primary contributors to heterogeneity in both the analysis of parity number Additionally, there are several important strengths of our study. First, to our knowledge, this is the first meta-analysis to report an association between maternal parity and CHDs. Moreover, our literature search was conducted on multiple databases, and the references from the retrieved articles were fully scrutinized to obtain any missing data. Therefore, our study included 43880 cases, enough to have sufficient statistical power to investigate the potential association between maternal parity and the risk of CHDs. Another strength of our study is that, although heterogeneity exists in our meta-analysis, we conducted a number of sensitivity, subgroup and Galbraith plot analyses and found that the results were stable.In summary, this study provides evidence that maternal parity number was positively associated with the risk of CHDs. However, more prospective studies, particularly in developing countries, are needed to further investigate the association between maternal parity and CHDs, especially with regard to the different subtypes of CHDs.Figure S1Galbraith plots for parity number (highest versus lowest) and CHD risk.(TIF)Click here for additional data file.Figure S2Galbraith plots for parity number (per 1 live birth) and CHD risk.(TIF)Click here for additional data file.Table S1Characteristics of studies of maternal parity and CHD risk.(DOC)Click here for additional data file.Checklist S1PRISMA checklist.(DOC)Click here for additional data file.


MTHFR) on childhood ALL risk in Taiwan, two well-known polymorphic genotypes of MTHFR, C677T (rs1801133) and A1298C (rs1801131), were analyzed to examine the extent of their associations with childhood ALL susceptibility and to discuss the MTHFR genotypic contribution to childhood ALL risk among different populations.Acute lymphoblastic leukemia (ALL) is the most prevalent type of pediatric cancer, the causes of which are likely to involve an interaction between genetic and environmental factors. To evaluate the effects of the genotypic polymorphisms in methylenetetrahydrofolate reductase . As for gender, the boys carrying the MTHFR C677T CT or TT genotype conferred a lower odds ratio of 0.51  for childhood ALL. As for age, those equal to or greater than 3.5 years of age at onset of disease carrying the MTHFR C677T CT or TT genotype were of lower risk .In total, 266 patients with childhood ALL and an equal number of non-cancer controls recruited were genotyped utilizing PCR-RFLP methodology. The MTHFR C677T T allele was a protective biomarker for childhood ALL in Taiwan, and the association was more significant in male patients and in patients 3.5 years of age or older at onset of disease.Our results indicated that the  The annMTHFR a potential candidate for a cancer-predisposition gene. It is reasonable that rapidly proliferating malignancies have a higher requirement for DNA synthesis and could be more susceptible to folate deficiency and resultant DNA damage. Low dietary folate and MTHFR deficiency induced the formation of intestinal tumors in a BALB/c mice model [To reveal the environmental and genomic factors together with the interactions among them is useful for evaluation and prevention of cancer risk. It is believed that one group of the candidate genes are those encoding enzymes related to the metabolism of identified carcinogens. Among these genes, methylenetetrahydrofolate reductase (MTHFR) is a folate-metabolism enzyme in charge of the conversion of 5, 10-methylene-tetrahydrofolate into 5-methyltetrahydrofolate, homocysteine remethylation, and biosynthesis of DNA and RNA . The regce model .MTHFR variations focused on the catalytic domain and the two polymorphisms, C677T (rs1801133) and A1298C (rs1801131), which may determine its enzymatic activity [MTHFR A1298C polymorphism is localized in the coding regulatory domain [MTHFR variant have found positive associations with solid cancers, such as colorectal cancer [Previous investigations of activity . In the 77T rs18033 and A177T rs18033 and A1y domain . As for y domain . In 20047T rs18013 and A12l cancer , breast l cancer , oral cal cancer , 17, andMTHFR genotypes had a significantly lower susceptibility to adult ALL and malignant lymphoma. In the same period, Franco [MTHFR C677T and A1298C in a representative pediatric population sample (control/case = 266/266), to investigate the correlation between MTHFR genotypes and childhood ALL in Taiwanese children, and to then summarize all of the relevant updated literature.Over the past decade, there has been a growing interest in the possible association between folate-related polymorphisms and the risk of developing lymphoid malignancies, including childhood ALL. First, Skibola  and Mats, Franco  and Wiem, Franco  providedOur study was approved by the Institutional Review Board of China Medical University Hospital, and written informed consent was obtained from one or both the parent of all participants. Two hundred and sixty-six patients diagnosed with childhood ALL  were recruited between Apr 2005 to Jan 2010 from the general surgery outpatient clinics within the Pediatric Departments at China Medical University Hospital and National Taiwan University Hospital, Taiwan, Republic of China. All of the clinical characteristics of these ALL patients, including their histological details, were identified by expert surgeons. All subjects voluntarily participated, completed a questionnaire with the help of parents or guardians and provided peripheral blood samples. The questionnaires recorded their disease history, diet and sleep lifestyles and the disease history, diet and behavioral lifestyle, social-economic status of the parents. An equal number of age-matched non-cancer healthy volunteers were selected for use as a control group following initial random sampling from the Health Examination Cohort established from Apr 2005 to Jan 2010 as previously published . The regMTHFR C677T and A1298C of all subjects was carried out by polymerase chain reaction restriction fragment length polymorphism (PCR-RFLP) assays as previously published [MTHFR C677T were forward 5’- TGA AGG AGA AGG TGT CTG CGG GA-3’ and reverse 5’- AGG ACG GTG CGG TGA GAG TG-3’. The primers for MTHFR A1298C were forward 5’- GGG AGG AGC TGA CCA GTG CAG-3’ and reverse 5’- GGG GTC AGG CCA GGG GCA G-3’. The underlined C mismatched base in the forward primer were used to create a partial Fnu4H I cutting site, which will be completed in the presence of the C allele at the A1298C polymorphic site. The following cycling conditions were performed: 5 min of initial denaturation at 95°C, 35 cycles of 30 sec of denaturation at 95°C, 30 sec of annealing at 54°C and 1 min of elongation at 72°C, and 7 min of final extension at 72°C. The 198-bp PCR product of MTHFR C677T and 138-bp PCR product of MTHFR A1298C were subject to enzyme digestion with Hinf I and Fnu4H I , respectively for 4 h and then visualized by ethidium bromide-stained 3% agarose gel electrophoresis under UV light. On digestion with Hinf I, the PCR product of MTHFR C677T arising from the C allele was uncut (198 bp), whereas the T allele was cut into fragments of 175 bp and 23 bp. On digestion with Fnu4H I, the PCR product of MTHFR A1298C arising from the A allele was uncut (138 bp), whereas the C allele was cut into fragments of 119 bp and 19 bp [Genomic DNA was prepared from peripheral blood leukocytes using a QIAamp Blood Mini Kit , long-term stored at -80°C, diluted and aliquotted for genotyping as working stock at -20°C –25. Genoublished , 25–27. P-value was less than 0.05. All statistical analysis were performed with SAS 9.2 and SPSS 17.Only those participants having both genotypic and clinical data (control/case = 266/266) were selected for final analysis. The descriptive statistics of patients and controls were presented as the mean and standard deviations (SDs) or as percentages. The Pearson’s chi-square test or Fisher’s exact test (when any cell was less than five) was used to compare the distribution of the genotypes. Associations were expressed and evaluated as odds ratios (ORs) with 95% confidence intervals (95%CIs). Interactions between the genotypes and onset age or gender were examined by using the likelihood ratio test. Statistical tests were deemed significant when the P>0.05) (The frequency distributions for the age and gender of 266 childhood ALL patients and 266 non-cancer controls are shown in P>0.05) .MTHFR C677T and A1298C in the controls and childhood ALL patients are shown in MTHFR SNPs of the controls were in Hardy-Weinberg Equilibrium . The genotypic frequency distributions for MTHFR C677T were significantly different between childhood ALL and control groups (P = 0.0076), while those for the A1298C polymorphism were not significantly different (P>0.05)  . Those wMTHFR genotype and age and gender were further analyzed and presented in rd and 134th subjects in the control and patient groups was 3.5 years; thus, we further stratified the groups into <3.5 and ≥3.5 year-old subgroups. Noticeably, in the elder (≥ 3.5 years) group, subjects with CT or TT genotypes for MTHFR C677T had lower risks for developing childhood ALL than those with the homozygous CC genotype ; however, this was not the case for the younger (<3.5 years) group , but this was not the case for the girls .Because age and gender are the predominant risk factors for developing childhood ALL, the interactions between the s) group . As for he girls . In summMTHFR gene may determine MTHFR enzyme activity and the availability of folate to the whole body and may ultimately be linked to childhood ALL risk. For this reason, we investigated their relationships with the susceptibility to developing childhood ALL in Taiwan. We found that the T variant genotypes of MTHFR C677T were significantly associated with a lower susceptibility to childhood ALL (MTHFR genotypes and childhood leukemia risk with a representative sample size (control/case larger than 70/70) and a non-redundant population. Among the studies, allele T was associated with decreased risk in populations of Taiwan, Serbia, China, Netherlands, Greece, Canada, and UK [MTHFR C677T and A1298C genotypes to childhood ALL, it was demonstrated that the T allele of MTHFR C677T was associated with a lower risk of childhood ALL in both Asians and Caucasians, while A1298C was not observed to have any effect [MTHFR C677T and negative for A1298C . After stratification by ethnicity, the significance only existed among Caucasians , at borderline among Asian , but not among others  [MTHFR genotypes to ALL susceptibility between boys and girls, further investigations are in urgent need. In the literature, studies have revealed significant differences in the serum or plasma folate levels among people with various C677T genotypes, but not among those with various A1298C genotypes [We have further analyzed the relationship between the C677T genotype and childhood ALL risk according to the subject age and gender. Interestingly, the interaction between  disease . Additioenotypes , 58. FurWe also tested for a gender-dependent effect on determining childhood susceptibility to ALL. The protective impact of the T allele at MTHFR C667T with respect to cancer risk appeared to be stronger for boys than the girls  but thisMTHFR and childhood ALL risk and investigated age- and gender-interactions with the genotype to determine childhood ALL susceptibility. The presence of the T allele of C677T was not only a detectable and predictive biomarker for childhood ALL but also a protective determinant for older patients and boys.In conclusion, this study documented the evidence of a relationship between the genotypes of


The previous published data on the association between the X-ray repair cross-conplementation group 1 (XRCC1) polymorphisms and thyroid cancer risk remained controversial. Hence, we performed a meta-analysis on all available studies that provided 1729 cases and 3774 controls (from 11 studies) for XRCC1 Arg399Gln, 1040 cases and 2487 controls for Arg194Trp (from 7 studies), and 1432 cases and 3356 controls for Arg280His (from 8 studies).PubMed, CNKI, and EMBASE database were searched to identify relevant studies. Overall, no significant association was found between XRCC1 Arg399Gln , Arg194Trp , and Arg280His  and thyroid cancer risk when all the eligible studies were pooled into the meta-analysis. In the further stratified and sensitivity analyses, significant association was still not found in these three genetic polymorphisms.In summary, this meta-analysis indicates that XRCC1 Arg399Gln, Arg280His, and Arg194Trp are not associated with thyroid cancer. Thyroid carcinomas are the most frequent endocrine malignancies which among these thyroid carcinomas, more than 90 percent are differentiated thyroid carcinomas (DTC). Pathologically, DTC include papillary, follicular, and Hürthle cell carcinoma XRCC (X-Ray cross-complementing) genes were initially discovered through their role in DNA damage response caused by ionizing radiation. They are important components of various DNA repair pathways contributing to DNA-damage processing and genetic stability XRCC1) is involved in the repair of DNA base damage and singlestrand DNA breaks by binding DNA ligase III at its carboxyl and DNA polymerase β and poly (ADP-ribose) polymerase at the site of damaged DNA XRCC1 have been identified at codon 194, 280, and 399  The XRCC1 polymorphisms at 194, 280, and 399  with thyroid cancer risk XRCC1 polymorphisms at 194, 280, and 399 with thyroid cancer risk XRCC1 Arg194Trp, Arg280His, and Arg399Gln with thyroid cancer risk have shown conflicting conclusions. In order to explore the association between Arg399Gln, Arg194Trp, and Arg280His polymorphisms with thyroid cancer risk, an updated meta-analysis was conducted to summarize the data. Meta-analysis is a good method for summarizing the different studies. It can not only overcome the problem of small size and inadequate statistical power of genetic studies of complex traits, but also provide more reliable results than a single case–control study.Many studies have reported the association of A bibliographical search was performed in PubMed, CNKI, and EMBASE database to identify studies that evaluated XRCC1 polymorphisms and thyroid cancer up to April 10, 2014. The search terms used were: (polymorphism or mutation or variant) and (XRCC1 or “X-ray repair cross-conplementation group 1”) and thyroid. The search was not limited to language. Additional studies were identified by hand searching references in original articles and review articles. Authors were contacted directly regarding crucial data not reported in original articles. In addition, studies were identified by a manual search of the reference lists of reviews and retrieved studies. We included all the case–control studies and cohort studies that investigated the association between XRCC1 Arg399Gln, Arg194Trp, and Arg280His polymorphisms and thyroid cancer risk with genotyping data. All eligible studies were retrieved, and their bibliographies were checked for other relevant publications. When the same sample was used in several publications, only the most complete information was included following careful examination.The included studies needed to have met the following criteria: (1) only the case–control studies or cohort studies were considered, (2) evaluated the XRCC1 Arg399Gln, Arg194Trp, and Arg280His polymorphisms and thyroid cancer risk, and (3) the genotype distribution of the polymorphisms in cases and controls were described in details and the results were expressed as odds ratio (OR) and corresponding 95% confidence interval (95% CI). Major reasons for exclusion of studies were as follows: (1) not for cancer research, (2) only case population, and (3) duplicate of previous publication.Information was carefully extracted from all eligible studies independently by two investigators according to the inclusion criteria listed above. The following data were collected from each study: first author's name, year of publication, country of origin, ethnicity, source of controls, genotyping method, and numbers of cases and controls in the XRCC1 Arg399Gln, Arg194Trp, and Arg280His genotypes whenever possible. Ethnicity was categorized as “Caucasian,” “African,” (including African Americans) and “Asian.” We considered the samples of studies from India and Pakistan as of “Indian’” ethnicity, and samples from Middle Eastern countries as “Middle Eastern” ethnicity. When one study did not state which ethnic groups was included or if it was impossible to separate participants according to phenotype, the sample was termed as “mixed population.” We did not define any minimum number of patients to include in this meta-analysis. Articles that reported different ethnic groups and different countries or locations, we considered them different study samples for each category cited above.versus Arg/Arg, Arg194Trp: Arg/Trp+Trp/Trp versus Arg/Arg, and Arg280His: Arg/Gln+His/His versus Arg/Arg); recessive model ; Homozygote model , Heterozygote model , and additive model , respectively. Heterogeneity assumption was checked by a chi-square-based Q test  I2 value, a value that describes the percentage of variation across studies that are due to heterogeneity rather than chance, where I2 = 0% indicates no observed heterogeneity, with 25% regarded as low, 50% as moderate, and 75% as high Q-statistic, which represents the magnitude of heterogeneity between-studies) P<0.05. Begg's funnel plots Crude odds ratios (ORs) together with their corresponding 95% CIs were used to assess the strength of association between the XRCC1 Arg399Gln, Arg194Trp, and Arg280His polymorphisms and thyroid cancer risk. The pooled ORs were performed for dominant model , 1,040 cases and 2,487 controls for Arg194Trp (from 7 studies), 1,432 cases and 3,356 controls for Arg280His (from 8 studies). –Relevant publications were retrieved and preliminarily screened. As shown in XRCC1 polymorphisms and thyroid cancer risk. For Arg399Gln, there was no significant association between this polymorphism and thyroid cancer risk in any genetic model when all the eligible studies were pooled together. Similarly, the combined results did not showed any association between Arg194Trp/Arg280His polymorphisms and thyroid cancer risk for all genetic models. However, in the subgroup analysis by ethnicity, the results showed that Arg/His genotype was associated with an increased risk of thyroid cancer among Caucasians . And carriers of the 399Gln variant allele have a decreased thyroid cancer risk in mixed population . We also detected that the Trp allele of Arg194Trp polymorphism was significantly increased thyroid cancer risk in mixed population . When subgroup analysis by histological subtype, the results showed that Arg194Trp polymorphism was associated with decreased papillary thyroid cancer (PTC) risk in dominant model .Ph = 0.089, Arg194Trp: Ph = 0.088, and Arg280His: Ph = 0.061), recessive model (Arg194Trp: Ph = 0.041), homozygote model comparison , heterozygote model (Arg280His: Ph = 0.035), and additive model comparison . Then, we assessed the source of heterogeneity by ethnicity and source of controls. The results of meta-regression indicated that ethnicity  but not source of controls  contributed to substantial heterogeneity among the meta-analysis. Although there were two studies There was significant heterogeneity among these studies for dominant model comparison  and PTC . For Arg280His polymorphism, when one study was excluded, the results were also changed in Caucasians .–P = 0.523 for dominant model, P = 0.466 for recessive model, P = 0.796 for additive model, P = 0.598 for homozygote model, and P = 0.329 for heterozygote model), Arg194Trp , and Arg280His , respectively.Both Begg's funnel plot and Egger's test were performed to access the publication bias of this meta-analysis. Begg's funnel plots did not reveal any evidence of obvious asymmetry in any genetic model in the overall meta-analysis –4. The r 70,000 protein XRCC1 polymorphisms and thyroid cancer risk, by critically reviewing 11 studies on XRCC1 Arg399Gln , 7 studies on Arg194Trp (1040 cases and 2487 controls), and 8 studies on Arg280His (1432 cases and 3356 controls).DNA is continuously damaged by endogenous and exogenous mutagens and carcinogens. The damages are fixed by multiple DNA repair pathways including base excision repair, nucleotide excision repair, mismatch repair, and double-strand break repair XRCC1 Arg399Gln, Arg280His, and Arg194Trp when all the eligible studies were pooled into the meta-analysis. And In the further stratified and sensitivity analyses, significant association was still not found in these three genetic polymorphisms. Zhu et al. Overall, no significant association was found between XRCC1 Arg399Gln, Arg194Trp, and Arg280His polymorphisms with thyroid cancer risk showed conflicting results. The study of Hu et al. Previous meta-analyses on There are several limitations in this meta-analysis. First, the controls were not uniformly defined. Although most of them were common populations, some controls were population-based; other controls were hospital-based. Hence, non–differential misclassification bias is possible. Second, in the subgroup analysis may have had insufficient statistical power to check an association, Third, we were also unable to examine the interactions among gene-environment, lacking of the original data of the included studies limited our further evaluation of potential interactions, which may be an important component of the association between XRCC1 Arg399Gln, Arg280His, and Arg194Trp polymorphisms and environment and thyroid cancer risk. Four, it was much difficult to get the all articles published in various language. Last, our results were based on unadjusted published estimates. Because of data limitations, we were unable to adjust them such as age and alcohol consumption et al. Our meta-analysis also has several strengths. First, a meta-analysis of the association of XRCC1 Arg399Gln, Arg280His, and Arg194Trp polymorphisms with thyroid cancer risk is statistically more powerful than any single study. Second, the quality of eligible studies included in current meta-analysis was satisfactory and met our inclusion criterion.In summary, this meta-analysis indicates that XRCC1 Arg399Gln, Arg280His, and Arg194Trp are not associated with thyroid cancer. However, it is necessary to conduct large sample studies using standardized unbiased genotyping methods, homogeneous cancer patients and well-matched controls. Moreover, further studies estimating the effect of gene–gene and gene–environment interactions may eventually lead to our better, comprehensive understanding of the association between the XRCC1 Arg399Gln, Arg280His, and Arg194Trp polymorphisms and thyroid cancer risk.Checklist S1PRISMA Checklist.(DOC)Click here for additional data file.


APOE) gene ε2/ε3/ε4 polymorphism with both cancer risk and circulating lipid profiles, aiming to examine the causal relevance between lipids and cancer risk. Article identification and data abstraction were conducted in duplicate and independently by two authors. Data were analyzed by STATA software. Twenty-five articles that examined the associations of APOE gene ε2/ε3/ε4 polymorphism with either cancer risk (n = 22) or circulating lipid changes (n = 4) were eligible. The presence of ε2 and ε4 alleles showed no overall associations with overall cancer risk when compared with ε3 allele. The ε4 allele was significantly associated with 1.40-fold  increased risk of developing cancer in Asian populations, and the presence of heterogeneity was low (I2 = 37.6%). Carriers of ε3/ε4 genotype had a significant reduction in circulating HDL-C  without heterogeneity (I2 = 16.6%). The predicted odds of having cancer for 1 mg/dL reduction in circulating HDL-C was 1.14 (95% CI: 1.00 to 1.89). The findings of this Mendelian randomization meta-analysis demonstrate that reduced circulating HDL-C might be a potentially causal risk factor for the development of overall cancer in Asians.We summarized published data on the associations of apolipoprotein E ( APOE genetic defectsAPOE  is polymorphic at two nucleotides, which yields 3 alleles  and 6 genotypes , with diverse receptor-binding capabilitiesAPOE gene ε2/ε3/ε4 polymorphism and cancer risk are conflicting and inconclusive578APOE gene ε2/ε3/ε4 genotypes, which would be of importance to provide background data to infer causality between circulating lipids and cancer risk. To fill this gap in knowledge and generate added information, we revisited this topic and summarized the associations of APOE gene ε2/ε3/ε4 polymorphism with both cancer risk and circulating lipid profiles in a large meta-analysis implementing Mendelian randomization technique.Some observational studies have revealed that people with low circulating cholesterol level tended to be more susceptible to many malignancies, such as lung cancer and breast cancer1APOE gene ε2/ε3/ε4 polymorphism with either cancer risk (n = 22) or circulating lipid changes (n = 4) were eligible according to the predefined inclusion and exclusion criteria5781011121314151617181920212324252627282930APOE-cancer association articles with 26 independent studies, there were 13478 cancer patients and 77592 controls in total. For 4 APOE-lipids association studies, data provided in both cancer patients and controls were analyzed separately, resulting in 6 independent studies for TG and 7 studies respectively for TC, HDL-C and LDL-C.Of 530 potentially relevant articles identified according to our search strategy, 25 articles that examined the associations of risk n =  or circuAPOE gene ε2/ε3/ε4 polymorphism with cancer risk and circulating lipid changes are presented in Baseline host characteristics of study populations for the associations of APOE-cancer association studies, breast cancer was reported in 10 studies, colorectal cancer in 8 studies, multiple cancers in 3 studies, prostate cancer in 2 studies, gastric, head and neck, hepatocellular cancers respectively in 1 study. 14 of 26 studies were conducted in White populations, and 4 respectively in Asian, Latinos and mixed populations. As for source of controls, 10 studies enrolled population-based controls, and 16 studies enrolled hospital-based studies. The majority of 26 studies were retrospective in design (n = 23) with the rest being prospective (n = 3). Cancer patients and controls were reported to be matched in 10 studies, unmatched in 10 studies and unreported in 6 studies. The mean age was significantly higher in cancer patients than in controls . No significance was observed in gender, BMI, smoking and family history of cancer between the two groups (P > 0.05).For 26 APOE gene ε2/ε2, ε2/ε4, ε4/ε4 genotypes, only allelic comparisons (ε2 versus ε3 and ε4 versus ε3) were computed. As shown in I2 = 20.3%), but significant heterogeneity for the comparison of ε4 with ε3 (I2 = 20.3%). The low probabilities of publication bias for both comparisons were reflected by the Begg's funnel plots , as compared with 8% reduced risk in Caucasian populations .In an attempt to examine whether risk prediction was heterogeneous between different subgroups, several subgroup analyses were conducted according to cancer type, ethnicity, source of controls, study design, matched status and sample size, respectively . There wAs age, gender, BMI, smoking and family history of cancer were continuous, several meta-regression models were constructed by including them as covariates separately, and still no significance was attained.APOE gene ε2/ε2, ε2/ε4, ε4/ε4 genotypes, mean lipid changes were only compared for genotype ε2/ε3 versus ε3/ε3 and ε3/ε4 and ε3/ε3 (I2 = 65.8%). In contrast, carriers of ε3/ε4 genotype had a significant reduction in circulating HDL-C  without heterogeneity (I2 = 16.6%). No statistical significance was observed for the other comparisons.In view of limited data on nd ε3/ε3 . Out of APOE gene ε4 allele and cancer risk in Asians and the relationship between ε3/ε4 genotype and circulating HDL-C reduction. The predicted odds of overall cancer for 1 mg/dL reduction in circulating HDL-C was 1.14 (95% CI: 1.00 to 1.89), and this estimate was significant at a significance level of 5% as the null hypothesis value of 1 was not included by the estimated 95% CI.At the requirements of Mendelian randomization technique, causal relevance between circulating lipids and cancer risk was only calculated based on the association between APOE gene ε2/ε3/ε4 polymorphism as a surrogate marker. This meta-analysis is unique to our knowledge, as it is to date the first to address the causal relevance between circulating lipids and cancer risk in medical literature.Extending the findings of a recent meta-analysis by Anand et al,APOE genetic alterations and circulating HDL profiles73536APOE gene ε4 allele was associated with significantly reduced HDL-C in circulation, reinforcing the soundness of selecting ε2/ε3/ε4 polymorphism as a surrogate marker. Besides, we observed that the ε4 allele was particularly overrepresented in Asian cancer patients relative to controls. Based on these observations, it is reasonably expected that low circulating HDL-C conferred by APOE gene ε4 allele is causally related with an increased risk of cancer in Asians. Nevertheless, given the inadequate statistical power of this meta-analysis in subgroup analyses, far larger sample sizes than examined here will be required to produce sufficient power to evaluate the causality between circulating HDL-C and cancer risk.Several observational and clinical studies have demonstrated an inverse association between circulating HDL-C and cancer risk; however, this association is currently subject to an ongoing debate, as the issues of confounding and reverse causation are intractable in classic epidemiology. Fortunately, Mendelian randomization has been introduced as a viable technique to overcome drawbacks of observational studies and obtain robust causal estimates4APOE gene ε2/ε3/ε4 polymorphism was selected in this study, and investigations on other candidate genes or polymorphisms involved in HDL-C regulation were highly encouraged, leaving a challengeable task to test whether this polymorphism integrated with other risk determinants will enhance cancer risk prediction. Fifthly, one key assumption of Mendelian randomization is that the genetic polymorphism under study should not exhibit a pleiotropic effect, which is beyond our capability in this meta-analysis to eliminate this effect. Nevertheless, the present meta-analysis enriched our understandings of circulating HDL-C in molecular carcinogenesis, which would facilitate the identification of at-risk individuals who would develop cancer later in future clinical screening.Several limitations of the present meta-analysis need to be acknowledged. Firstly, we restricted our search scope to published articles written in only English language, and we cannot totally rule out the likelihood of selective publication bias. Secondly, almost all involved studies had circulating lipids measured only once, which cannot reflect its long-term profile in the development of cancer. Thirdly, this meta-analysis was based on summarized data, rather than individual participant data, precluding further gene-to-environment interactions. Fourthly, only Taken together, the findings of this Mendelian randomization meta-analysis demonstrate that reduced circulating HDL-C might be a potentially causal risk factor for the development of overall cancer in Asians. For practical reasons, it is encouraging to deem this study as a beginning instead of an endpoint of investigations to establish and optimize the background data to understanding the causal relevance of circulating HDL-C to carcinogenesis of multiple solid tumors.The present meta-analysis was carried out in accordance with the guidelines formulated in the PRISMA  statement (see the Supporting Checklist)APOE gene ε2/ε3/ε4 polymorphism with cancer risk or circulating lipid changes, we systematically searched PubMed and Embase electronic databases as of December 20, 2014 using the following subject terms, ‘apolipoprotein E or apo E or APOE or apo-E′, in combination with ‘cancer or carcinoma or neoplasia or tumor or adenoma or neoplasm or myeloma or melanoma or lymphoma or leukaemia or leiomyoma’ and ‘polymorphism or variant or variation or mutation or genotype or allele or SNP’. We also manually checked the reference lists of major original articles and reviews for the missing citations of relevance.To identify all relevant articles that assessed the associations of The titles and abstracts of all retrieved articles were independently read by two authors of this meta-analysis (Chunhua Yang and Xuri Li) to assess their eligibility. If we cannot reject an article with certainty, its full text was reviewed to ascertain whether relevant data were provided and if necessary we contacted study authors by emails to request additional information. We extracted data from the most recent or complete article if a same study group was reported by more than one publications. This process was run in duplicate and independently by the same two authors, and any uncertainty over the eligibility was adjudicated by a discussion or further joint inspection of original articles.APOE gene ε2/ε3/ε4 polymorphism and all sites of cancer except for skin were provided; (b) regarding circulating lipid changes, the mean or medium values and the corresponding standard deviation of circulating lipids including triglyceride (TG), total cholesterol (TC), high-density lipoprotein cholesterol (HDL-C) and low-density lipoprotein cholesterol (LDL-C) were provided across APOE gene ε2/ε3/ε4 alleles or genotypes; (c) study design should be either prospective or retrospective; (d) detailed genotype or allele counts of APOE gene ε2/ε3/ε4 polymorphism were tractable between cancer patients and controls.All studies that met the following criteria were included: (a) regarding cancer risk, data on associations between APOE gene ε2/ε3/ε4 polymorphism or that were lack of cancer-free controls.Conference abstracts or proceedings that did not specifically address the topic of our analysis were excluded from full-text review. Case reports or series, editorials, narrative or systematic reviews, or non-English articles were also not covered. Also this meta-analysis did not involve studies that examined the progression, severity or response to treatment or survival of cancer in association with APOE gene ε2/ε3/ε4 polymorphism between cancer patients and controls, the mean or medium (standard deviation) values of circulating TG, TC, HDL-C and LDL-C for each APOE gene ε2/ε3/ε4 allele or genotype carriers, as well as baseline characteristics of study populations when available such as age, gender, body mass index (BMI), smoking and the family history of cancer. The units of circulating TG, TC, HDL-C and LDL-C were uniformly standardized as mg/dL for consistency.Data were gathered independently from each eligible article by two authors (Chunhua Yang and Xuri Li) according to a predefined protocol developed by all contributing authors, including the first author's last name, publication year, ethnicity, cancer subtype, case-control matched status, source of controls, study design, sample size, the genotype and/or allele counts of All statistical analyses were managed with the use of STATA software  on Windows.APOE gene ε2 or ε4 allele with cancer risk was expressed as odds ratio (OR) and 95% confidence interval (95% CI) when compared with the ε3 allele. Considering the confounding effect of heterogeneity between studies, only random-effects model with the DerSimonian & Laird methodThe association of The probability of publication bias was assessed by visual Begg's funnel plot and the Egger's test, as well as the trim-and-fill method which can infer the existence of unpublished hidden articles from a filled funnel plot and correct the meta-analysis by imputing the presence of missing studies to yield an unbiased pooled estimate.I2) statistic, which ranges from 0% to 100% and is defined as the percentage of the observed between-study variability that is due to heterogeneity rather than chance. In this meta-analysis, I2 > 50% is designated as a threshold to indicate significant heterogeneityHeterogeneity was quantified by the inconsistency index (To evaluate the impact of individual studies on pooled effect estimates, we performed sensitivity analysis by sequentially omitting each study one at a time and computing differential estimates for remaining studies.APOE gene ε2/ε3/ε4 polymorphism and cancer risk to that of the relationship between this polymorphism and circulating lipid changes to reflect the possible causal relevance of these lipids on cancer.Under the assumptions of Mendelian randomization as formulated by Katan MB in 1986G.T., J.M. and B.W. conceived and designed the experiments; C.Y. and XuriL performed the experiments; C.Y. and G.T. analyzed the data; X.W., XuriL, XianglinL and W.W. contributed materials/analysis tools; C.Y., G.T. and B.W. wrote and revised the manuscript. All authors reviewed and approved the manuscript prior to submission.Supplementary materials


Cervical cancer (CC) has one of the highest mortality rates among women worldwide. Several efforts have been made to identify the genetic susceptibility factors underlying CC development. However, only a few polymorphisms have shown consistency among studies.We conducted a systematic review of all recent case-control studies focused on the evaluation of single nucleotide polymorphisms (SNPs) and CC risk, stringently following the “PRISMA” statement recommendations. The MEDLINE data base was used for the search. A total of 100 case-control studies were included in the meta-analysis. Polymorphisms that had more than two reports were meta-analyzed by fixed or random models according to the heterogeneity presented among studies.p21 rs1801270 polymorphism (C/A+A/A) and CC . We also found a negative association with the rs2048718 BRIP1 polymorphism dominant inheritance model (T/C+C/C) and CC , as well as with the rs11079454 BRIP1 polymorphism recessive inheritance model and CC . Interestingly, we observed a strong tendency of the meta-analyzed studies to be of Asiatic origin (67%). We also found a significant low representation of African populations (4%).We found significant negative association between the dominant inheritance model of p21 rs1801270 polymorphism, as well as BRIP1 rs2048718 and rs11079454 polymorphisms, with CC risk. This study suggests the urgent need for more replication studies focused on GWAS identified CC susceptibility variants, in order to reveal the most informative genetic susceptibility markers for CC across different populations.Our results provide evidence of the negative association of  Even though it is the cancer with the greatest demonstrated potential for secondary prevention, cervical cancer (CC) constitutes a significant public health problem. It has the third highest cancer incidence rate in women worldwide and even the second highest in some developing countries, like Mexico ,2. In 20According to records from Globocan 2012 and assuming that the estimated CC incidence rates remained constant in Mexico, an alarming 28% increase in CC cases in this region is estimated by the year 2020 . GloballAlthough several factors that contribute to CC development have been identified—mainly intrinsic factors (genetic), and extrinsic factors belonging to the Human Papillomavirus (HPV)—genetic factors show great potential for use as susceptibility or prognosis factors ,9. Most Therefore, we formulated the following PICO  question: Which SNPs reported in literature confer susceptibility or protection to CC in the international consensus? To address it we conducted a systematic review and meta-analyses following “PRISMA” statement recommendations  14]..14].A systematic literature search led by the PICO question was conducted to identify articles covering SNPs associated with the risk of developing cervical cancer (CC). From this question, a search in the MEDLINE database was performed through the PubMed database browser with the combination of the following terms: “polymorphism”, “single nucleotide polymorphism”, “SNP”, “uterine cervical neoplasm”, “cervical intraepithelial neoplasia” and “cervical cancer” from November 2009 to February 2015. The search was limited to the studies published 2009 onwards to ensure homogenization in the genotyping technique used across studies. The search was performed with the restriction of language of the full-text to English, Spanish or Portuguese. The scope of computerized literature search was expanded according to the reference lists of retrieved articles. The original articles were screened manually by two independent reviewers (TPK and MNGA). If the full text of an article was not available online to perform the screening we proceed to contact the first author of the article by e-mail requesting a copy of their work.Studies concerning the association of SNPs with the risk of developing CC were included if the following conditions were met: (i) any study describing the association of SNPs with CC; (ii) genotypes or allelic frequencies reported for control and case groups, given that for the pooled ORs calculation this data is strictly needed; (iii) case-control or nested case-control studies, since other epidemiological designs might result in heterogeneous case definitions; (iv) any study that reported the numbers of both controls and CC cases; (v) include at least 100 cases, in order to include only does studies with the minimal number of cases necessary to have a representation of all three genotypes ; (vii) not focused on HLA antigen genetic markers. The latter selection criteria is due to the high concordance among studies over HLA haplotypes association with CC, and the aim of the present work is to evaluate the association of CC with genetic polymorphisms that remains to be fully accepted as genetic susceptibility factors for the disease (which is not the case of HLA studied haplotypes).in situ, invasive cervical cancer, cervical adenocarcinoma or cervix cancer). Those studies evaluating only the association between precancerous lesions  were excluded, as precancerous lesions do not always lead to cancer development. In the case that a study assessed the association with both precancerous lesions and cervical cancer, only the group of cases defined as histologically confirm CC were taken into account, if the study did not differentiate between precancerous lesion and CC patients the study was excluded. This criteria also applied for those studies assessing other types of cancer besides CC in the same report.Since our event of interest was CC risk all selected studies should defined as main outcome CC at any stage ; study design; number of controls and cases; genotyping method; distribution of age, gene name, polymorphism and genotypic frequency. The data was extracted and registered into two databases by two reviewers (TPK and MNGA) independently. The reviewers were blind to journal names, institutions or fund grants. Any discrepancy between these two expert investigators in the field was resolved by a third reviewer (MMV), who also extracted the data and participated in the discussion with them and made the final decision.Statistical analyses were performed for each gene polymorphism for which at least two studies were available. First, Odds Ratios (OR), with their respective confidence intervals (CI), were estimated by individual study based on the primary data obtained from the review.2 type tests for heterogeneity between and intra-study, known as Cochran’s Q, were performed. The null hypothesis for this test asserts that the association is the same in each study and therefore the observed variability is explained by chance alone [2 was estimated [2 was between 25 and 50%, moderate if the I2 was between 50 and 75%, and high if the I2 was greater than 75% [2 test was also applied, whose null hypothesis and interpretation is equivalent to the Cochran’s Q.Pooled ORs were estimated for each polymorphism and Χce alone ; in contstimated , which ithan 75% . In ordeQ or the Τ2 were also considered and not only the I2. In that sense, if any of the two former measures yielded a significant result with an α of 0.05 , then random effects models were applied. In the case in which both tests were not statistically significant  fixed effects models were used.As a stringent approach to decide if the random effects model, rather than the fixed effects model, was used for considering the weights in the pooled OR estimation for each polymorphism, the afore mentioned Cochran’s Sensibility analyzes were carried out for the polymorphisms with more than four studies by iteratively removing a study to repeatedly recalculate the pooled OR, and finding consistency in all cases. Finally, potential publication bias was evaluated with an Egger Test only in polymorphisms with more than five studies regardless of whether significant association was found in the meta-analysis ,18.For all the analyses a statistical significance α level of 0.05 was considered. The α level was set at this value as this is the most widely used in the majority of the original studies of the field. Furthermore, the definition of the significance level was made taking into consideration the limitation of the statistical power of the meta-analysis performed for each association in order to avoid a possible type 2 error . NeverthAll analyses were performed in STATA v.14 .In the search for eligible studies, we inputted the aforementioned keywords into MEDLINE and obtained a total of 1115 studies. When we applied the filters for language and year of publication we ended up with 419 studies to be screened. Only 118 studies met the criteria specified in the method section to be further assessed in full-text. Of these, 22 articles were excluded since they did not report genotypic frequencies or had a sample size of less than 100 CC cases . We werep53 rs1042522 polymorphism was the most reported with seven studies in total . All the meta-analyzed SNPs were reported in at least three different studies, with 42.86% SNPs reported in more than three studies. The in total .For each SNP, the pooled OR and 95% CI for the heterozygous genotype, minor allele homozygous genotype, dominant inheritance model , recessive inheritance model , and for the minor allele were calculated. In all models the heterogeneity between studies was taken into account when selecting which meta-analysis model to apply (random effects or fixed effects model).rs2048718 BRIP1 gene polymorphism   (2 = 0.00 and Tau2 = 0.00) so the fixed effects model was used   and the  = 0.03) . Both ofwas used .A/A) carriers for the rs1801270 p21 gene polymorphism and CC risk   carriers and CC risk did not reach statistical significance   . However = 0.82) .rs1801270 p21 gene polymorphism maintained the significant negative association previously observed for the heterozygous and the minor allele homozygous carriers when assessing the association by dominant  (p<0.01) inheritance models  (BRIP1 gene polymorphism (rs11079454) showing a p value of 0.04 (The  p<0.01)  and recee models . Additio = 0.03) . Althoug of 0.04 .rs1801270 p21 gene polymorphism minor allele  had a statistically significant association with CC risk   we could not detect significant association with CC risk showing a p value of 0.12 and 0.19, respectively  , which wectively .Given that the number of studies per polymorphism was so limited, we could not perform meta-analyses stratifying by ethnicity. More than half (67%) of the studies included in the review were from an Asiatic population, 17% from a Caucasian population and only 4% from an African population. In fact, the median of the number of studies included in the review was of two studies per country, with an interquartile range of three studies. The country with most studies included was China with 44 studies, followed by India with 11 studies and Poland with 10 studies .p53 rs1042522 polymorphism had the minimum five studies required to assess the possibility of publication bias. The Egger test for the heterozygous, minor allele homozygous and allele models are presented as funnel-plots in Only the p21 gene polymorphism (rs1801270), two BRIP1 gene polymorphisms (rs2048718 and rs11079454) and CC. Carriers of one or two copies of the minor allele  of the p21 gene polymorphism (rs1801270) were 32% less likely to develop CC than carriers of the homozygous genotype for the ancestral allele (C/C). Likewise, carriers of at least one copy of the rs2048718 BRIP1 gene polymorphism minor allele  were 20% less likely to develop CC than carriers of the homozygous genotype for the ancestral allele (T/T). We only found association with the recessive inheritance model for the rs11079454 polymorphism of this gene, in which the carriers of two copies of the minor allele (A/A) were 27% less likely to develop CC than carriers of one or two ancestral alleles (T/A + T/T).The main findings of our study were a significant negative association between the et al, 2014 [et al, 2015 [To our knowledge, to date only two previous studies of extensive meta-analysis of genetic polymorphisms and CC have been made ,21. Unlial, 2014 . Furtheret al, 2015 [et al, 2014 [et al, 2015[p21 gene polymorphism (rs1801270) which we consistently reported as a protective factor for CC in our study was not found to be significantly associated with CC by any of these authors.Since there was a limited number of studies by polymorphism, it was not possible to analyze ethnic subgroups, as performed by Wang al, 2015 . In contal, 2014  and Wang al, 2015, did notp21 gene  encodes a 21-kda protein and was first described as a potent inhibitor of cell proliferation and DNA replication, both in physiological conditions and after DNA damage [p53, p21 functions as a unique link between p53, cell-cycle arrest and DNA repair, exhibiting anti-oncogenic properties [The A damage . As the operties ,24.Atm-deficient mice results in a delay in thymic lymphomagenesis and in an apparent increase in spontaneous apoptosis in tumor cells [On the other hand, studies focused on the role of p21 in ATM-mediated signal pathways have shown that p21 is a downstream effector of ATM-mediated growth control, and have suggested that loss pf p21 in or cells ,26. Thisor cells .p21 polymorphisms affect protein expression and activity, and hence might play a role in susceptibility to cancer [p21 rs1801270 (Ser31Arg) polymorphism and the risk of several types of cancers , though the results remain inconclusive [Several studies have suggested that o cancer ,28. In tnclusive ,29–31 dup21 rs1801270 polymorphism produces a C to A transversion and causes a substitution from serine (Ser) to arginine (Arg), affecting the DNA binding zinc finger domain of the protein [A allele of this polymorphism does not affect the previously reported tumor suppressor activity of p21 [A allele  in CC is still uncertain. protein . Interesy of p21 . Howeverp21 rs1801270 polymorphism and gastrointestinal tract tumor risk reported that the ancestral allele  might be a risk factor for this type of cancer among Asiatic women [et al. [et al. [C allele) might be a risk for CC. Although in previous meta-analyses centered on CC risk this polymorphism was not reported as significantly associated, we believe that the environmental and gene interactions have an important role over the effect of this SNP over CC development, given the complexity and duality of its role over oncogenesis.A previous meta-analysis focused only on the ic women . Furtheric women . In that [et al.  and Qiu  [et al. , since wBRIP1 gene codifies a DNA-dependent ATPase and a DNA helicase denominated BRCA1-interacting protein 1, which belongs to the RecQ DEAH helicase family [BRIP1 gene has been associated with other types of cancer, especially breast cancer [The e family . Since it cancer –38.BRIP1 was found to be overexpressed in advanced squamous cervical cancer biopsies of non-responsive to base-line therapy patients, as compared to biopsies of responsive patients[BRIP1 gene in the risk of CC [Particularly for CC,  patients. Howeversk of CC ,41.BRIP1 polymorphisms we found to be associated with CC risk fall in the untranslated region (UTR) of the gene. The rs2048718 BRIP1 polymorphism is located in the 5´-UTR, whereas the rs11079454 polymorphism is in the 3´-UTR.Both of the rs2048718 or rs11079454 BRIP1 polymorphisms have in CC development, there is evidence suggesting that at least rs11079454 might be affecting the expression of the BRIP1 gene [rs11079454 is in LD (linkage disequilibrium) with other three BRIP1 SNP , and the haplotype of this LD block including the ancestral allele  of rs11079454 polymorphism was associated with CC risk among Chinese women, which is consistent with the association observed in this meta-analysis [Assessing the functionality of polymorphisms located in the UTR is quite a challenge and, even though there are no reports of the direct effect IP1 gene ,41. The analysis .p53 rs1042522 polymorphism. The shortage of studies per polymorphism could be explained by the fact that GWAS (Genome-wide association studies) increasingly reveal new variants associated with CC, but replication studies in different populations for these variants are scarce. It is, therefore, difficult to conclude the true role of such variants in CC susceptibility. There is a clear need for replication studies that can validate the association of the GWAS discovered genetic variants and reveal their role in genetic susceptibility to CC in each population. Similarly, studies from several candidate genes on their own do not provide conclusive and reproducible data for other populations, despite reporting statistically significant associations. Finally, we cannot guarantee that population stratification did not affect the results of the constituent studies in the meta-analyses, since we could not perform an analysis stratified by ethnicity.The meta-analysis at hand has limitations, and the interpretation of the results presented here should be made in the context of these. In the first instance, to avoid the potential bias of genotyping inaccuracy given by the genotyping method used in the studies, we included only recent studies which limited the number of studies per polymorphism. The reduced number of reports per SNP included affected the statistical power of the study to detect small effects. Additionally, this affected the assessment of potential publication bias, as it was only possible to evaluate it in the From a wider perspective, lack of reproducibility of the candidate SNP studies is unfortunately a general trend beyond CC. The boom of GWAS generated a global effort to study a large number of SNP without a clear path nor a previous biological hypothesis, leading to a picking and fishing approach. This practice was acceptable in the beginning of the genomic age but nowadays a systematic effort that guarantees reproducibility is highly required .p21 rs1801270 polymorphism, as well as BRIP1 rs2048718 and rs11079454 polymorphisms, and CC risk. Additionally, this study highlights the urgent need for more replication studies focused on GWAS identified CC susceptibility variants.Notwithstanding the presence of heterogeneity due to factors not addressed in this study, our results provide evidence of the negative association between S1 File(PDF)Click here for additional data file.S2 File(DOCX)Click here for additional data file.S1 Table(XLS)Click here for additional data file.S2 Table(XLSX)Click here for additional data file.S3 Table(XLSX)Click here for additional data file.