Breast (BCa) and gynecological (GCa) cancers constitute a group of female neoplasms that has a worldwide significant contribution to cancer morbidity and mortality. Evidence suggests that polymorphisms influencing miRNA function can provide useful information towards predicting the risk of female neoplasms. Inconsistent findings in the literature should be detected and resolved to facilitate the genetic screening of miRNA polymorphisms, even during childhood or adolescence, and their use as predictors of future malignancies. This study represents a comprehensive systematic review and meta-analysis of the association between miRNA polymorphisms and the risk of female neoplasms. Meta-analysis was performed by pooling odds-ratios (ORs) and generalized ORs while using a random-effects model for 15 miRNA polymorphisms. The results suggest that miR-146a rs2910164 is implicated in the susceptibility to GCa. Moreover, miR-196a2 rs11614913-T had a moderate protective effect against female neoplasms, especially GCa, in Asians but not in Caucasians. MiR-27a rs895819-G might pose a protective effect against BCa among Caucasians. MiR-499 rs3746444-C may slightly increase the risk of female neoplasms, especially BCa. MiR-124 rs531564-G may be associated with a lower risk of female neoplasms. The current evidences do not support the association of the remaining polymorphisms and the risk of female neoplasms. Breast (BCa) and gynecological (GCa) cancers constitute a group of female neoplasms ICD-10: C50, C51–58) that has significant contribution to cancer morbidity and mortality worldwide. Despite advances in the screening and treatment of cancer ,3, BCa i0, C51–58miRNAs are short non-coding RNAs that are involved in negatively regulating the expression of most protein-coding genes in the post-transcriptional level . They arThis study represents a comprehensive systematic review and meta-analysis of the risk of female neoplasms that are associated with miRNA polymorphisms. We were interested in finding: which miRNA polymorphisms are hypothesized to modify the risk of female neoplasms, including breast and gynecological cancers in the literature, whether there are sufficient data to draw robust conclusions regarding the role of miRNA polymorphisms in the susceptibility to different female neoplasms, whether miRNA polymorphisms pose ethnic-specific effects in the susceptibility to female neoplasms. n: 1779), review articles (n: 1018), abstracts or conference papers (n: 495), studying other genes or polymorphisms (n: 179), studying other diseases (n: 138), systematic reviews or meta-analyses (n: 122), and not written in English (n: 20). The full texts of the remaining 79 records were evaluated and 10 articles were excluded for the following reasons: insufficient reported data (n: 4) . However, the results of pooling effect sizes under genetic models showed no significant association , Caucasians (n: 3), Africans (n: 1), or South Americans (n: 1). Moreover, most studies (n: 14) evaluated the risk of BCa and only a few studies (n: 4) focused on GCa. In four studies, the genotype distributions of the control group significantly deviated from the HWE expectations [P < 0.01], which indicates that, with the C allele being the mutant and the T allele being the reference, the mutational load of miR-499 rs3746444 is implicated in increased susceptibility to female neoplasms. The pooling of effect sizes under genetic models revealed that rs3746444 was associated with the risk of female neoplasms while assuming the homozygote (CC vs. TT) and the allelic models (C vs. T)  , suggestIn the sensitivity analysis, excluding the low-quality studies led to a significant diminution of the heterogeneity in all modes of inheritance, especially the homozygote and the recessive models . ExcludiMYH7B gene .Six studies containing 2211 cases and 2422 controls were included in the meta-analysis of the association between miR-149 rs2292832 and the risk of female neoplasms ,68,80,87ctations ,65. Metan: 3), Africans (n: 1), or South Americans (n: 1). All of the studies explored the association between miR-605 rs2043556 and BCa. In three studies, the genotype distributions of the control group significantly deviated from the HWE expectations [Five studies containing 2706 cases and 3804 controls were included in the meta-analysis ,38,55,87ctations ,30,55. Mctations , it shoun: 4) or South Americans (n: 1). All of the studies explored the association between miR-608 rs4919510 and BCa. Meta-analysis using the ORG and the OR under the different genetic models showed that miR-608 rs4919510 was not associated with BCa in all studies , Africa-Americans (n: 1), or European-Americans (n: 1). Four studies explored the association of miR-100 rs1834306 and BCa and two studies evaluated GCa. According to the quality scores . Excluding the low-quality study did not influence the conclusion and the results remained significant under the homozygote and the recessive models , Caucasians (n: 1), and African-Americans (n: 1). Three studies explored the association between miR-34b/c rs4938723 and BCa and one study evaluated GCa . A meta-analysis using the ORG revealed no evidence for an association between miR-34b/c rs4938723 and female neoplasms [2: 60.30). No significant association was observed when the three studies on BCa were pooled , sufficient data for performing a meta-analysis were not available and, therefore, no conclusion regarding the contribution of these polymorphisms to the risk of female neoplasms was drawn in the current study. The following conclusions may be obtained based on the results of this study concerning 15 miRNA polymorphisms that were included in the meta-analysis. For most miRNA polymorphisms, breast cancer was the most studied cancer, followed by the cervical or ovarian cancer, among studies that were included in the meta-analysis. Only a few studies were focused on other types of gynecological cancers and different types of gynecological cancers were less studied than breast cancer. Moreover, it should be noted that this study did not adjust for covariates, like age and gender or interaction with environmental factors, and it was based on unadjusted summary effects of the original studies. Designing a flexible platform for data harmonization might provide the feasibility of performing meta-analysis of individual patient data, which allows for the testing of interactions with covariates [In conclusion, this study systematically identified 65 miRNA polymorphisms that were evaluated for possible contribution to the risk of female neoplasms . For the majority of studied polymorphisms . The following keywords or MeSH terms were used.  AND (miRNA OR microRNA OR pre-mir OR miR) AND (“Single nucleotide polymorphism” OR SNP OR variant OR variation OR polymorphism OR mutation OR locus). References of previous meta-analysis, review articles, or other relevant articles were also manually screened to identify potentially eligible articles. This meta-analysis was carried out in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement .Studies meeting all of the following criteria were included: (1) a case-control study evaluating the association between a polymorphism in any miRNA gene and susceptibility to female neoplasms , including breast cancer (ICD-10: C50) and tumors of the female genital tract; (2) availability of the genotype count data for estimating the odds ratio (OR) and its 95% confidence interval (95%CI); (3) the full-text article is written in English; and, (4) a minimum of two studies per each miRNA polymorphism should be met to include the polymorphism in the meta-analysis. Studies that met any of the following criteria were excluded: (1) meta-analysis, review articles or abstracts; (2) duplicate publications; (3) studies on animals or cell-lines; (4) studies without a case-control design; (5) studies that did not report genotype counts or allele frequencies; (6) studies investigating survival, progression, or severity of the disease; and, (7) the article was not written in English.Each eligible study was screened and the following items were recorded: the first author, publication year, country, ethnicity, the cancer type, the source of controls, the miRNA name, polymorphism ID, genotyping method, genotype counts for each SNP, and number of cases and controls recruited. The quality of each study was assessed while using a modified version of quality assessment criteria for genetic association studies used elsewhere ,118,119 2 were used to assess the significance of heterogeneity and the Z test was performed to assess the significance of pooled ORs . Considerable heterogeneity was expected between the studies because of the differences in sources of the control group and other sample characteristics. Therefore, the random-effects model (RE) with the DerSimionian–Laird estimator was used to calculate the summary effects in all cases [Associations of miRNA polymorphisms with female neoplasms are represented with pooling odds ratios (ORs) and their 95% confidence intervals (95% CIs) assuming five genetic models . Moreover, the generalized odds ratio (ORG), which is a model-free measure of genetic risk effect, was also used to evaluate the association between miRNA polymorphisms and female neoplasms . The ORGll cases . In the ll cases .2, was below 0.1 . In the presence of substantial heterogeneity , the method that was proposed by Rücker et al. [Univariate meta-regression was carried out to identify potential sources of heterogeneity. Subgroup analyses that were based on the cancer type, ethnicity, and the study quality were also performed. When examining and interpreting the asymmetry of funnel plots, the recommendations of other investigators have been followed . As statr et al. , an arcs2 when each study is removed in turn, the test statistics for  heterogeneity when each study is removed in turn, the diagonal elements of the hat matrix, and the weights (in %) that are given to the observed outcomes during the model fitting [http://biomath.med.uth.gr) [Genotype frequencies of polymorphisms in the control group of each study were assessed for the departure from HWE expectations while using the exact goodness of fit test. The following approach was employed to deal with studies in which the genotype counts in the control groups deviated from HWE expectations . HWE-deviated studies were not excluded from the meta-analyses. However, a sensitivity analysis was performed to evaluate the possible influence of excluding HWE-deviated studies on the pooled ORG and corresponding 95% CIs. When more than ten studies were included in the meta-analysis, influential study diagnostics for the model-free approach were computed using the metafor package for R . The fol fitting ,125. The fitting . Cook′s  fitting . The cov fitting . A study fitting ,125: (i).uth.gr) . All oth.uth.gr) .


Objectives The uridine diphosphate glucuronosyltransferase 1A1 (UGT1A1)*28 allele in HIV-positive patients receiving atazanavir (ATV) might be associated with the risk of hyperbilirubinemia. Owing to mixed and inconclusive results, a meta-analysis was conducted to systematically summarize and clarify this association.Methods Based on a comprehensive search of PubMed, Embase and Web of Science databases, studies investigating the association between UGT1A1 alleles and hyperbilirubinemia was retrieved. We evaluated the strength of this relationship using odds ratios (ORs) with 95% confidence intervals (CIs). Sensitivity analysis was performed by removing each study one at a time and calculating the pooled ORs of the remaining studies to test the robustness of the meta-analysis results. The Q statistic and the 2I index statistic were used to assess heterogeneity. Publication bias was evaluated using Orwin’s fail-safe N test.Results A total of six individual studies were included in this meta-analysis. A significantly increased risk of hyperbilirubinemia was observed in HIV-positive patients receiving ATV with the UGT1A1*1/*28 or UGT1A1*28/*28 genotype, and the risk was higher with the UGT1A1*28/*28 genotype than with the UGT1A1*1/*28 genotype. . All of the pooled ORs were not significantly affected by the remaining studies and different modeling methods, indicating robust results.Conclusions This meta-analysis suggests that the UGT1A1*28 allele represents a biomarker for an increased risk of hyperbilirubinemia in HIV-positive patients receiving ATV. Atazanavir (ATV) is currently one of the most widely used protease inhibitors in the treatment of HIV infection. It displays several advantages over other protease inhibitors, such as a favorable lipid profile, once daily dosing, low capsule burden, and a relatively distinct resistance profile . The mosUGT1A1 is involved in the glucuronidation of several commonly used drugs, including gemfibrozil, ezetimibe, simvastatin, atorvastatin, cerivastatin, ethinylestradiol, buprenorphine, ibuprofen and ketoprofen . An outsThere is some clinical controversy over the relevance of ATV-associated hyperbilirubinemia; it is benign and reversible upon discontinuation of the drug. Thus, some believe a proactive test in an attempt to predict and prevent its occurrence may not be warranted because they consider this adverse effect is not clinically relevant , whereasTwo independent investigators (P.D. and A.W.) carried out a systematic search in PubMed, Embase and Web of Sciences databases, with the last search update on 20 February 2019. The following search terms were used: (‘UGT1A1’ OR ‘UGT1A1*28’) AND ‘ATV’ AND (‘unconjugated hyperbilirubinemia’ OR ‘hyperbilirubinemia’), without any limitation applied. The reference lists of retrieved studies and recent reviews were also manually searched for further relevant studies.Hyperbilirubinemia refers to the accumulation of bilirubin in the blood causing jaundice. The clinical manifestations of jaundice are the skin, sclera and mucous membranes appear yellow. Studies in this meta-analysis must meet the following inclusion criteria: (i) evaluation of the association between UGT1A1*28 and hyperbilirubinemia; (ii) case–control study or cohort study; (iii) studies focusing on humans; and (iv) detailed genotype data in HIV-positive patients. Exclusion criteria: (i) duplication of previous publications; (ii) comment, review and editorial; and (iii) study with no detailed genotype data. Study selection was achieved by two investigators independently according to the inclusion and exclusion criteria by screening the title, abstract and full text. Any dispute was solved by discussion.The data of the eligible studies were extracted in duplicate by two investigators independently (P.D. and X.L.). The following contents were collected: PMID, first author name, year of publication, country of origin, the characteristics of cases and controls, ethnicity, total bilirubin levels, sample size, Hardy–Weinberg equilibrium, number of cases and controls, study type and Newcastle–Ottawa Scale (NOS). The two authors checked the extracted data and reached a consensus on all data. If a dissent existed, they would recheck the original data of the included studies and have a discussion to reach a consensus. The authors used a quality score, ranging from 0 to 9, with higher scores associated with better study quality. Two investigators scored the studies independently and solved disagreements through discussion using the NOS Scoring system .2I statistic was calculated to quantitate the proportion of the total variation across studies due to heterogeneity [P<0.1 or I2 > 50%) among the studies. The overall effects were calculated by a combined Z value, with P<0.05 indicating statistical significance.Our analysis was conducted using the Review Manager Software 5.3 . A chi-square-based Q test was used to measure heterogeneity between studies. In addition, an ogeneity . Fixed-eOdds ratios (OR) and 95% confidence intervals (95%CI) were calculated to evaluate the strength of the association between the UGT1A1*28 allele and its susceptibility to hyperbilirubinemia. The UGT1A1*28 allele contained UGT1A1 *1/*28 and *28/*28, and the common genotype was UGT1A1*1/*1. For the present study, we used the following group: (i) UGT1A1*28/*28 versus UGT1A1*1/*28; (ii) UGT1A1*1/*28 versus UGT1A1*1/*1; (iii) UGT1A1*28/*28 versus UGT1A1*1/*1; (iv) UGT1A1*28/*28 versus UGT1A1(*1/*28 + *1/*1); and (v) UGT1A1(*1/*28 + *28/*28) versus UGT1A1*1/*1.2I values. If 2I > 50% or P<0.1, we considered the study to have a great degree of heterogeneity and used a random-effects model. Otherwise, a fixed-effects model would be more appropriate.Sensitivity analysis was conducted by the removal of each study, one at a time, understanding the effect of exclusion of respective studies by calculating the pooled ORs of the remaining studies. Heterogeneity was measured using 0.05) for each meta-analysis, was applied to assess the publication bias. As a rule of thumb, if the Nfs0.05 was higher, the authors of the meta-analysis could be more comfortable assuming that the sample of studies was not likely to be overwhelmed by a future influx of studies with no significant relationship [0.05 was calculated according to the formula:In addition, the fail-safe number, with significance set at 0.05 . Eighty-eight articles were excluded, of which 43 were duplicate articles and 45 had no relation to this topic. The remaining 14 studies were full-text reviewed, and 8 studies were excluded. Among these excluded studies were one having no relation to hyperbilirubinemia ,26, one P=0.0003). Heterogeneity was not statistically significant across all studies   A.2I > 50%. Four included trials compared the risk of hyperbilirubinemia between HIV-positive patients with a UGT1A1*1/*28 genotype and those with a wild-type allele [2I = 61%, P=0.05). Overall analyses suggested an increased risk of hyperbilirubinemia in HIV-positive patients with a UGT1A1*1/*28 genotype as compared with those with a wild-type allele   B.P<0.00001). No statistical heterogeneity was detected across all studies   C.P<0.00001). No statistical heterogeneity was detected across all studies   D.P<0.00001). Moderate heterogeneity was detected across all studies ; and five analyses combined subgroup also suggested an increased risk of hyperbilirubinemia in HIV-positive patients with UGT1A1*28 allele as compared with those with a wild-type allele . Large heterogeneity was detected across all studies   E.0.05 = 2.50; (2) UGT1A1*1/*28 versus UGT1A1*1/*1: Nfs0.05 = 11.97; (3) UGT1A1*28/*28 versus UGT1A1*1/*1: Nfs0.05 = 23.94; (4) UGT1A1 (*1/*28 + *28/*28) versus UGT1A1*1/*1: Nfs0.05 = 48.03; and (v) UGT1A1*28/*28 versus UGT1A1 (*1/*28 + *1/*1): Nfs0.05 = 31.09. Based on these comprehensive quantitative evaluations, we concluded that no obvious publication bias existed in the assessed research.All of the pooled ORs were not significantly affected by the remaining studies and different modeling methods, indicating robust results . MoreoveP=0.001) in the remaining 24 patients in the switch arm, 12 months after therapy simplification, leading to a significant reduction in the number of patients presenting ≥ grade 3 hyperbilirubinemia. That means the combination of ritonavir and ATV increased the plasma concentration of ATV, which resulted in a higher bilirubin level than ATV alone. Bioequivalence between cobicistat and ritonavir as a pharmacoenhancer of ATV was established. Cobicistat shows increased advantages over ritonavir, such as no activity against HIV, fewer drug–drug interactions and better solubility, which promotes coformulation strategies with less pill burden, better tolerability, and, potentially, higher life-long treatment adherence [To the best of our knowledge, this is the first meta-analysis examining the effect of UGT1A1*28 polymorphism on the risk of hyperbilirubinemia in HIV-positive patients receiving ATV. The unconjugated bilirubin level was calculated by subtracting the conjugated bilirubin level from the total bilirubin level. Hyperbilirubinemia severity was classified on the basis of the AIDS Clinical Trial Group guidelines  for totadherence . Lankiscdherence  clarifieP=0.01). Culley et al. [It is important to note that four additional important articles were not included in this meta-analysis . The letter  and the y et al.  confirmey et al.  found thUGT1A1 gene [UGT1A1 gene as opposed to six (TA6) that characterizes the wild-type allele (UGT1A1*1) [UGT1A1 is a major conjugating enzyme responsible for bilirubin homeostasis , and it 1A1 gene . Some haGT1A1*1) . The disGT1A1*1) ,46. GilbGT1A1*1) . The *36GT1A1*1) . The UGTGT1A1*1) .Polymorphisms in UGT1A1 are associated with indirect bilirubin concentrations in the general population (i.e. Gilbert’s syndrome). Polymorphisms in genes beyond UGT1A1 have been reported to be associated with serum bilirubin concentrations in the general population, including ABCC2, ABCB4, ABCB11, ATP8B1, SLCO1B1 , SLCO1B3Limitations of this meta-analysis must be considered. First, the possibility of information and selection biases cannot be completely excluded because some of the included studies were retrospective. Second, we restricted our search to articles published in English or Chinese. Articles with potentially high-quality data that were published in other languages were not included because of anticipated difficulties in obtaining accurate medical translation. Third, our study did not make the correlation analysis of ethnicity and drug doses. Finally, the association of hyperbilirubinemia and ATV mainly occurs when ATV is boosted. Hypothetical selection bias could have selected patients all with boosted ATV, and that this association might not exist in non-boosted ATV regimens.In conclusion, the presence of the UGT1A1*28 allele with ATV use increases the risk of developing severe hyperbilirubinemia. Although hyperbilirubinemia is considered a mild adverse effect, it has clinical implications. Jaundice causes discomfort due to the yellowish appearance of the skin, which may affect the quality of life of these patients and may lead to treatment discontinuation. It is important to keep in mind that the variant allele frequencies should be considered in each population before initiating a genotyping program.


CHD) in offspring is uncertain; additionally, a complete overview is missing. A meta‐analysis of observational studies was performed to address the question of whether women who had a history of viral infection in early pregnancy were at an increased risk of CHD in offspring, compared with mothers without viral infection.At present, the association between maternal viral infection and risk of congenital heart diseases  and random‐effects models  suggested that mothers who had a history of viral infection in early pregnancy experienced a significantly increased risk of developing CHD in offspring. For specific viral infections, the risk of developing CHD in offspring was significantly increased among mothers with rubella virus  and cytomegalovirus  in early pregnancy; however, other maternal viral infections in early pregnancy were not significantly associated with risk of CHD in offspring. Sensitivity analysis yielded consistent results. No evidence of publication bias was observed.Unrestricted searches were conducted, with an end date parameter of July 15, 2018, of PubMed, Embase, Google Scholar, Cochrane Libraries, and Chinese databases, to identify studies that met prestated inclusion criteria. Seventeen case‐control studies involving 67 233 women were included for analysis. Both fixed‐effects models (odds ratio [OR], 1.83; 95% CHD in offspring.Although the role of potential bias and evidence of heterogeneity should be carefully evaluated, the present study suggests that maternal viral infection is significantly associated with risk of  Today, the association between maternal viral infection and risk of congenital heart disease in offspring is still uncertain; a complete review of the evidence to date has been lacking.Our study represents the first meta‐analysis of risk of congenital heart disease associated with maternal viral infections.Our study shows that maternal viral infection, in particular infection with rubella or cytomegalovirus, in early pregnancy is significantly associated with risk of congenital heart disease in offspring, which indicates that early detection and intervention for viral infection may help to reduce the incidence of congenital heart disease.Congenital heart disease (CHD) is the most common congenital birth defect. CHD accounted for nearly one third of all birth defects, and the prevalence rate reached to 8 to 12 per 1000 live births worldwide.Despite recent advances in medicine and surgery, CHD continues to cause much perinatal morbidity and mortality. It is reported that CHD caused ≈303 300 deaths in 2015, which was decreased from the 366 000 since 1990.Several studies have demonstrated that a number of genetic and environmental factors have been associated with the development of CHD in the fetus. Furthermore, some risk factors have been identified, such as phenylketonuria, rubella, retinoic acid, and the use of certain specific drugs.However, available evidence to date has not found any consistent association between maternal viral infections and risk of CHD, and there are many conflicting results across previous studies. For example, some studies indicated that there is a weak or null association between maternal viral infections and CHD.The authors declare that all supporting data are available within the article.We attempted to report the present systematic review and meta‐analysis by following the recommendations of the Preferred Reporting Items for Systematic Reviews and Meta‐analyses statement.In this review, the exposures of interest were maternal viral infections. Mothers who had a history of viral infection, such as rubella virus, cytomegalovirus, hepatitis B virus, coxsackievirus, and herpesvirus, were defined as the exposed group, and those without viral infection as the unexposed group. The outcomes of interest were CHDs. We first performed an initial screening of titles or abstracts. A second screening was based on full‐text review. Studies were considered eligible if they (1) were published in Chinese or English; (2) had a cohort or case‐control design; (3) had use of maternal viral infection as the exposure of interest; (4) had use of CHDs as outcomes of interest; and (5) reported relative risks and odd ratios (ORs), with corresponding 95% CIs .Data extraction was performed using a standardized data collection form. Any reported relative risks or ORs and their 95% CIs of CHD for mothers who had a history of viral infection compared with those without viral infection, as well as characteristics for each study, were extracted. Additionally, any reported relative risks or ORs and their 95% CIs describing the relationship between different types of viral infection and risk of CHD were also extracted. Corresponding information was recorded as follows: the first author's name, year of publication, geographic region, study design type, sample size, diagnostic methods of CHD, ascertainment of viral infection, viral infection time, reported types of viral infection, reported ORs and their 95% CIs, whether the confounding factors were adjusted when estimating the association between maternal viral infection and risk of CHD, and quality score.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp) to assess the quality of included studies. In statistics, the scale is a tool used for assessing the quality of nonrandomized studies included in a meta‐analysis. Using the tool, each study is judged on 8 items, categorized into 3 groups: the selection of the study groups; the comparability of the groups; and the ascertainment of outcome of interest for cohort studies. Stars awarded for each quality item serve as a quick visual assessment. Stars are awarded such that the highest‐quality studies can be awarded as many as 9 stars. When the study gains ≥6 stars, it is considered of higher methodologic quality.Two authors (Z.W.Y. and J.B.Q.) independently conducted the studies selection, data extraction, and quality assessment. Any disagreements were resolved through discussion among the authors until consensus was reached. We adopted the principles of the Newcastle–Ottawa Scale .Statistical analysis was performed using Stata version 12.0  and Review Manager version 5.3 . Statistical tests were declared significant for a 2‐sided P<0.10), and the Egger linear regression test (significance level at P<0.10). Subgroup analyses, sensitivity analyses, and publication bias assessment were performed only for the association between maternal total viral infection and risk of CHD in offspring because of rather small numbers of studies for specific viral infections.Subgroup analyses were conducted according to study design type, ascertainment of viral infection, diagnostic method of CHD, whether confounding factors were controlled, and quality score. We further conducted a sensitivity analysis to examine the influence of various exclusion criteria on the overall risk estimate. We also investigated the influence of a single study on the overall risk estimate by omitting 1 study in each turn. Potential publication bias was assessed by using the Begg's funnel plots, the Begg's rank correlation test  and random‐effects models  suggested that mothers who had a history of viral infection in early pregnancy experienced a significantly increased risk of developing CHD in offspring when compared with those without viral infection  was observed across studies.Both fixed‐effects models  and cytomegalovirus  in early pregnancy compared with the reference group. However, we did not find a significant association for other maternal viral infections . However, other maternal viral infections in early pregnancy were not significantly associated with risk of CHD in offspring , cytomegalovirus , and other virus .Random‐effects models suggested that mothers infected with rubella virus in early pregnancy were at a significantly higher risk of developing CHD in offspring  was identified as the most relevant heterogeneity moderators. When data were restricted to adjusted studies , the risk of developing CHDs increased further. However, there was no statistically significant difference for risk of CHD associated with viral infection between adjusted data and unadjusted data . Additionally, differences for risk of CHD associated with viral infection were not statistically significant between the remaining subgroups . Overall, mothers with viral infection were still at higher risk of CHD in offspring among most subgroup data except for lower‐quality studies.Subgroup analyses for the association between maternal total viral infection and risk of developing CHD in offspring were summarized in Table P<0.001; I2=89%). Exclusion of 13 studies in which the quality scores were <8 yielded similar results , with substantial evidence of heterogeneity . Further exclusion of any single study did not materially alter the overall risk estimates , yet heterogeneity was still present  and the Egger linear regression test  also indicated no evidence of publication bias among studies of maternal viral infection and CHD risk in offspring.Visual inspection of the Begg's funnel plot did not identify substantial asymmetry Figure . FurtherCHDs constitute the most common birth defects among newborns and have emerged as one of the most important causes of infant mortality. Additionally, CHDs have a significant impact on child and adult morbidity and disability.2 range, 40%–83%).Findings from the present meta‐analysis indicated that mothers infected with a virus in early pregnancy were at a significantly increased risk of developing CHD in offspring when compared with those without viral infection . When data were restricted to adjusted studies , the risk of developing CHD increased further. For specific viral infections, our study showed that mothers with rubella virus and cytomegalovirus in early pregnancy had a higher risk of developing CHD in offspring compared with mothers without viral infections. However, our review did not indicate that other maternal viral infections, including herpes virus, hepatitis B virus, and coxsackievirus, significantly increased the risk CHD in offspring. Of note, these results must be viewed with caution because of some evidence of heterogeneity ; all included studies were considered to be of higher methodologic quality. All of the original studies included focused on the link between maternal viral infection in early pregnancy and CHD, which, because the critical period of heart development is in early pregnancy,Although our study suggested that the risk of CHD in offspring was significantly increased among mothers with viral infections, the reasons are not clear, which was rarely discussed in previous studies. Previous studiesHeterogeneity is often a concern in a meta‐analysis. In the present review, substantial heterogeneity was found among studies of maternal viral infections and CHD risk, which was not surprising given the differences in study population characteristics, study design types, ascertainment of viral infection, diagnostic method of CHD, adjustment for confounding factors, and study quality. We used subgroup analyses to explore the potential sources of heterogeneity on the basis of the above‐mentioned characteristics. As a result, whether the confounding factors were adjusted was identified as the most relevant heterogeneity moderator. So far, there has been a general consensus that CHD is a polygenic disease caused by both genetic and environmental factors. Many factors, including maternal socioeconomic situation, education, dietary pattern, diabetes mellitus, obesity, metabolic syndrome, alcohol consumption, smoking, and genetic background, have been reported to be associated with risk of CHD.Potential limitations of this study should be considered in future studies. First, in our review, we estimated the risk of total CHD associated with maternal viral infections, but we did not estimate the risk of specific CHD subtypes. In fact, during the design phase of our study, we did consider estimating the risk of specific CHD subtypes associated with maternal viral infections. However, all of the studies included did not provide data on specific subtypes of CHD. CHDs include many pathological types, such as atrial septal defects, ventricular septal defects, patent ductus arteriosus, tetralogy of Fallot, and so on. Future studies should focus on the association between maternal viral infection and specific CHD subtypes.Second, although the present review did include studies published in Chinese or English, all included studies were from China, except one study from Canada, which may restrain the popularization and application of present findings. After repeated confirmation, our search strategy was correct. Therefore, additional research in other populations is warranted to generalize the findings.Third, residual confounding is a concern. Uncontrolled or unmeasured risk factors have the potential to produce biases. Although restricting analysis to studies that have matched or adjusted confounding factors did not materially alter the combined risk estimate, the possibility cannot be ruled out that residual confounding affected the results because these factors do not explain all CHD risk.Fourth, estimates were substantially heterogeneous across studies. Nevertheless, we were able to detect the major source of heterogeneity through the subgroup analysis and the sensitivity analysis. The sensitivity analysis that omitted one study at a time and calculated the combined OR for the remaining studies yielded consistent results. After subgroup analysis, the heterogeneity was obviously decreased. However, our estimates must be viewed with caution because of heterogeneity.Fifth, the assessment of maternal viral infections and CHD diagnosis were different across studies, which may lead to classification bias. For example, in all included studies, 9 studies (52.9%) used laboratory testing to assess whether mothers were infected with a virus; while the remaining studies (47.1%) used questionnaires. Therefore, we cannot rule out the possibility that the mother's viral infection may be underreported. In addition, all included studies are case‐control studies, which may lead to recall bias. Furthermore, for risk estimates of CHD associated with specific viral infections, the results mainly relied on between 3 and 7 of the 17 total studies, so more studies should be included in future reviews to provide further support for our results. Last but not least, after reviewing the funnel plot, we can identify 4 of 17 papers outside the funnel, which means that around 25% of studies included had important publication bias. Therefore, potential publication bias could influence the findings.Our study, which includes a large number of participants, giving it sufficient statistical power, aimed to address the question of whether mothers who were infected with a virus in early pregnancy were at a higher risk of CHD in offspring compared with those without viral infections. Although the role of potential bias and evidence of heterogeneity should be carefully evaluated, our study indicated that maternal viral infections in early pregnancy are significantly associated with risk of CHD in offspring. However, the underlying mechanisms involved in the association between maternal viral infections and CHD risk are uncertain and require further study for elucidation. In the future, some large and multicenter prospective cohort studies will need to confirm our findings.Qin was supported by the Project Funded by Natural Science Foundation of Hunan Province (2018JJ2551), National Natural Science Foundation Program of China (81803313), Hunan Provincial Key Research and Development Program (2018SK2063), and New Teachers’ Scientific Research Driven Foundation of Central South University (502045001). The sponsor of the study had no role in study design, data collection, data analysis, data interpretation, or writing of the report.None.


Overall hazard ratio for progression-free survival was 0.58 , with significant heterogeneity and publication bias, and unbiased estimate was significant . Overall hazard ratio for overall survival was 0.51 , and this estimate was not likely confounded by heterogeneity or publication bias. Subgroup and meta-regression analyses suggested that hypertension grade of controls, sample size, age and gender were possible causes of heterogeneity. Taken together, our findings indicate that bevacizumab-induced hypertension can predict progress-free survival and overall survival in patients with metastatic colorectal cancer, whereas its prediction for objective response rate was nonsignificant.We tested the hypothesis that bevacizumab-induced hypertension may be a useful predictor for objective response rate, progression-free and overall survival in patients with metastatic colorectal cancer via a comprehensive meta-analysis. Search process, article selection and data extraction were independently performed by two investigators. Statistical analyses were conducted using the STATA/SE software. Fourteen independent studies and 2292 study subjects were synthesized. Overall relative risk of objective response rate for bevacizumab-induced hypertension was 2.03 (95% confidence interval [CI]: 1.18-3.48,  Colorectal cancer is the third most common cancer in the world, and its global burden is projected to be increased by 60% to over 2.2 million new cases and 11 million deaths by 2030 . It is eBevacizumab, a humanized anti-vascular endothelial growth factor (VEGF) monoclonal antibody, is increasingly recognized as standard of care for the treatment of metastatic colorectal cancer in a first-line setting, with reasonable biological implications ,11. HoweIn total, 632 articles were identified after searching the medical literature, as well as the reference lists of retrieved major articles and reviewers. Of them, only 13 articles including 14 independent studies and 2292 study subjects were qualified for the final analysis -15,18-26The baseline characteristics of 14 qualified studies published from 2009 to 2016 are listed in p=0.01), whereas this risk was clouded by the significance of between-study heterogeneity (I2: 77.1%).As shown in p<0.001) and was not impacted by heterogeneity (I2: 0.0%), with the effect estimate over twice as much as that in studies with hypertension grade 0 in controls . Moreover, no significance was detected after dividing studies using the median cutoff value of total sample size at 77, in spite of divergent difference in risk estimates. In univariate meta-regression analysis, gender was identified as a significant source of heterogeneity for overall response rate (p=0.037).Heterogeneity sources were explored using both subgroup analysis and meta-regression analysis. In subgroup analysis, bevacizumab dose, sample size and hypertension grade of controls may account for the presence of significant heterogeneity based on significant differences in stratified relative risk . For exap=0.047). Analysis of incorporating the three missing studies showed that the unbiased relative risk of response rate was 1.55, which did not deviate significantly from 1 . Further cumulative meta-analysis indicated a stable trend in the risk estimates of objective response rate for bevacizumab-induced hypertension, as presented in Publication bias was evaluated using both filled funnel plots  and Eggep<0.001), with significant heterogeneity (I2: 74.7%)  .I2: 0.0%). In meta-regression analysis, age and gender were identified as significant confounders for the prediction of bevacizumab-induced hypertension for progression-free survival .Subgroup analysis showed that hypertension grade of controls and sample size may confound the prediction of bevacizumab-induced hypertension in patients with metastatic colorectal cancer based on the differences in hazard ratio between subgroups . The risp=0.001). After adjusting for the five missing studies, the hazard ratio of progression-free survival was still significant . In cumulative meta-analysis, a stable trend in risk estimates was noted for progression-free survival , and this estimate was not likely confounded by heterogeneity (I2: 23.1%), as presented in p<0.001) and with total sample size over the median cutoff value of total sample size at 84 . Meta-regression analysis showed that age and gender might be other causes of heterogeneity .The pooled hazard ratio for overall survival of bevacizumab-induced hypertension was 0.51 . Risk estimates were stabilized in cumulative meta-analysis, as shown in No missing study was reported by filled funnel plot in Finally, trial sequential analysis was employed to minimize random errors for objective response rate, progression-free survival and overall survival, respectively . The thrThe aim of this present study was to update the results of two previous meta-analyses ,17 by inIt is widely recognized that VEGF is a key mediator of angiogenesis and an effective biological target for patients with metastatic colorectal cancer, and its overexpression can accelerate tumor progression and metastatic spread of colorectal cancer ,28. AlsoPreviously, two meta-analyses examined the validity of using hypertension to predict response rate and survival of patients with metastatic colorectal cancer, and both studies consistently demonstrated that bevacizumab-induced hypertension was associated with significant improvement in objective response rate, progression-free survival and overall survival ,17. DiffBesides objective response rate, we have assessed the association of bevacizumab-induced hypertension with two survival outcomes. Consistent with the results of two previous studies, our findings consolidated the prognostic contribution of bevacizumab-induced hypertension to both progression-free survival and overall survival. However, extending the results of the two studies, we found that sample size and hypertension grade of controls, as well as age and gender, may be possible causes of between-study heterogeneity. In particular, after grouping studies per the median cutoff point of total sample size, the prediction of bevacizumab-induced hypertension for survival outcomes was more obvious when analysis was restricted to the large studies, indicating the robustness of our findings. In fact, the difference between progression-free survival and overall survival is whether the incorporation of patients who get worse. Just because of this difference, another aspect worth noting was that risk magnitude of overall survival was more obvious in studies with hypertension grade 0 of controls than studies with hypertension grade 1/2, whereas this situation was slightly reversed for that of progression-free survival. In other words, the protective effect of bevacizumab indexed by overall survival was more obvious for a lower grade of hypertension, whereas that by progression-free survival was more obvious for a higher grade of hypertension. Generally, the probability of tumor aggressiveness is higher during early treatment of bevacizumab for metastatic colorectal cancer than the late treatment. A higher grade of bevacizumab-induced hypertension may surrogate a longer treatment period. So the findings of this study further demonstrated the usefulness of using hypertension to signal bevacizumab treatment in metastatic colorectal cancer. Moreover, it is not surprising to notice that the prediction of bevacizumab-induced hypertension for survival outcomes was age- and gender-dependent. As evidenced, aging-related methylation can influence the gene expression of key control genes in colorectal cancer and adenoma , and horSome limitations deserve special considerations for this meta-analysis. Firstly, only articles published in the English language were identified and the exclusion of gray literature from this meta-analysis can lead to the exaggerated estimates of intervention effectiveness . SecondlTaken together, through a comprehensive analysis of 14 independent studies and 2292 study subjects, our findings indicate that bevacizumab-induced hypertension can predict progress-free survival and overall survival in patients with metastatic colorectal cancer, whereas its prediction for objective response rate was nonsignificant. Meanwhile, several issues from this meta-analysis remain to be clarified, and yet other areas warrant further investigation. If the usefulness of bevacizumab-induced hypertension were successfully validated in the future, it will be clinically important to administrate bevacizumab agents to patients with metastatic colorectal cancer who are more likely to benefit from the treatment.The conduct of this meta-analysis was consistent with the guidelines in the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) statement .A systematic literature search was conducted in the following electronic bibliographic databases: PubMed (Medline), EMBASE (Excerpta Medica dataBASE) and Web of Science . Search strategy was expressed in the Boolean style, that is,  and (“cancer” or “tumor” or “tumour” or “carcinoma” or “neoplasm”) and (“bevacizumab” or “avastin”) and (“hypertension” or “blood pressure”). The literature search was completed on February 17, 2018. The reference lists of two previous meta-analyses ,17 and rArticles were retained for analysis pending the simultaneous satisfaction of following criteria: (i) metastatic colorectal cancer patients receiving bevacizumab treatment were grouped according to the presence or severity of hypertension determined by the Common Terminology Criteria for Adverse Events (CTC AE); (ii) information on objective response rate or its associated odds ratio, or hazard ratio for progression-free survival or overall survival, or survival curves was available for extraction or inference; (iii) articles were written in the English language. Meanwhile, conference abstract, case report or case series, review or meta-analysis, basic medical research and phase I clinical trial were not included in this study.Two investigators (Chun-Jing Zhang and Shu-Ying Zhang) of this present study independently assessed the eligibility of all potential articles according to above criteria. A third investigator (Hai-Tao Yu) solved the disagreements, if exist, from literature screening.The following variables were extracted from each eligible article: the first author’s surname, published year, country where study was conducted, CTC AE version, combined chemotherapy, bevacizumab dose, age, gender, metastatic position of metastatic colorectal cancer, objective response rate, progression-free survival, overall survival, odds ratio (95% CI) and hazard ratio (95% CI). Survival rate was used to estimate progression-free survival or overall survival from survival curves in case of no available risk estimates by aid of the Engauge Digitizer software Release 4.0.Data extraction was independently completed by two investigators (Chun-Jing Zhang and Shu-Ying Zhang) of the present study, and a third investigator (Hai-Tao Yu) checked table entries for accuracy by referring to original context.According to a previous publication , trial sOdds ratio and its 95% CI for objective response rate, as well as hazard ratio and its 95% CI for progression-free survival or overall survival were calculated between metastatic colorectal cancer patients with and without bevacizumab-induced hypertension.From a statistical standpoint, in the absence of between-study heterogeneity, effect estimates based on the fixed-effects model and the random-effects model are exactly the same, whereas in the presence of heterogeneity, effect estimates are more reliable based on the random-effects model, relative to the fixed-effects model. Thus, in this meta-analysis, random-effects model using the DerSimonian & Laird method ,41 was uI2) on the basis of the Cochrane Q-test. Heterogeneity is reported to be low if I2 ranges from 0% to 25%, moderate from 25% to 75% and high from 75% to 100% [Between-study heterogeneity was quantified using the inconsistency index ( to 100% .Regardless of the magnitude of heterogeneity, subgroup analysis and meta-regression analysis were conducted to assess the impact of categorical and continuous study variables on the overall heterogeneity.The presence of publication bias, the bias resulting from the possibly unpublished studies due to the negative results or extreme deviations from previous results, was visually appraised by filled funnel plots. The statistical evidence of publication bias was assessed by the Egger’s test, a weighted regression test that can help justify the asymmetry of funnel plots. In case of evident publication bias, a trim and fill method was employed to derive an unbiased estimate after considering potentially missing studies. In addition, to evaluate whether later studies impacted previous studies, a cumulative meta-analysis was conducted accordingly.Above statistical analyses were completed using the STATA/SE software Release 11.2.Supplementary File


Background: Findings by epidemiologic studies on menopausal hormone replacement therapy (HRT) and the risk of ovarian cancer are inconsistent. This study aimed to assess the association of menopausal HRT with the risk of ovarian cancer by histological subtype.Methods: A literature search was performed in PubMed, Web of Science, and EmBase for relevant articles published from inception to August 2018. Pooled relative risk ratios (RRs) with 95% confidence intervals (CIs) were determined with a random-effects model.Results: Thirty-six studies involving 4, 229, 061 participants were included in this meta-analysis. The pooled RR of ovarian cancer was 1.29  for menopausal HRT. In subgroup analysis by study design, pooled RRs of ovarian cancer in cohort and case-control studies were 1.35 (95%CI 1.19–1.53) and 1.24 (95%CI 1.11–1.38), respectively. In subgroup analysis by continent, association of menopausal HRT with ovarian cancer was significant for North America (1.41 [1.23–1.61]), Europe (1.22 [1.12–1.34]), and Asia (1.76 [1.09–2.85]), but not Australia (0.96 [0.57–1.61]). Association differed across histological subtypes. Increased risk was only found for two common types, including serous (1.50 [1.35–1.68]) and endometrioid (1.48 [1.13–1.94]) tumors.Conclusion: This meta-analysis suggests that menopausal HRT may increase the risk of ovarian cancer, especially for serous and endometrioid tumors. Ovarian cancer is known as the most lethal genital system malignancy . It is aMenopausal hormone replacement therapy (HRT) is widely used to improve postmenopausal symptoms and ward off bone loss. However, in the past few years, many epidemiological studies have revealed that HRT is associated with an increased risk of breast cancer , 10. DatWe performed a literature search to identify relevant available articles from PubMed, Web of Science and EmBase from inception to August 2018 with no restrictions. Search terms included “hormone replacement therapy” (or “HRT”) and “ovarian cancer” (or “ovarian neoplasms” or “ovarian carcinoma” or “ovary cancer”). The reference lists of the included studies were also reviewed for potential relevant studies.Inclusion criteria were: (1) original report from observational studies; (2) menopausal HRT as the exposure of interest; (3) ovarian cancer as the outcome of interest; (4) relative risk ratio (RR) with 95% confidence interval (CI) provided. The most recent and complete study was selected if studies from the same population were repeated.Two investigators searched and reviewed all relevant studies independently. Any disagreement was resolved by consensus with the involvement of a third reviewer.The following information were extracted from each study by two investigators independently: first author's name, published year, country, study design, follow-up duration, age range or mean age at baseline, sample size and number of cases, histological subtype of ovarian cancer, the types of hormones used in the study population, RR  with 95%CI and adjustment for potential confounders. We extracted RRs adjusted for the most confounding factors in the original studies. We prioritized the RRs for highest vs. lowest duration category of HRT use. If the study did not provide RRs for highest vs. lowest duration category of HRT use, we extracted the RRs for “use vs. non-use.”I2 statistic was adopted to assess heterogeneity among studies . Meta-regression with restricted maximum likelihood estimation was performed to explore the important covariates that might have significant impact on between-study heterogeneity. Subgroup analyses were stratified on study design, geographic location and the types of hormones used in the study population. Sensitivity analysis was performed with one study removed at a time to assess whether the results could have been affected markedly by a single study. The funnel plot and Egger's test were performed to explore the small-study effect.The Newcastle–Ottawa Scale was used to assess the quality of studies included in this meta-analysis. Pooled data were obtained as the inverse variance-weighted means of the logarithm of RRs with 95%CI to assess the associations of menopausal HRT and the risk of different histological subtypes of ovarian cancer, respectively. The DerSimonian and Laird random effects model (REM) was used to combine study-specific RRs (95%CIs). The P-values) were two-sided, with a statistical significance level of 0.05.All statistical analyses were performed with STATA version 14.0 . All reported probabilities  and 1.24 , respectively , Europe , and Asia , but not Australia   and 1.85 , respectively  . In subgectively .The associations of menopausal HRT with the risk of ovarian cancer in various histological subtypes are summarized in I2 = 27.5%, Pheterogeneity = 0.175) and endometrioid  tumors. In subgroup analysis stratified by study design, pooled RRs for serous tumors in cohort and case-control studies were 1.64  and 1.42 , respectively. In subgroup analysis stratified by continent, significant positive associations were found for North America  and Europe , respectively. In subgroup analysis stratified by the hormones types, pooled RRs for HRT and ERT were 1.48  and 1.54 , respectively. In subgroup analysis for endometrioid tumors, significant positive associations were obtained in cohort studies  and Europe , respectively.In the five major histologic subtypes, significant positive associations were observed in serous , and the pooled RR was 1.27 (95%CI 1.17–1.37).To assess between-study heterogeneity, we performed univariate meta-regression with the covariates of study design, publication year and continent. However, none of these covariates was found to have a significant impact on between-study heterogeneity. After excluding two study , 20  of the association of menopausal HRT with the risk of ovarian cancer ranged from 1.27 (95%CI 1.18–1.38) to 1.31 (95%CI 1.20–1.42). No individual study had excessive effect on the pooled RR.Povarian cancer = 0.083) showed no evidence of significant small-study effect for the association of menopausal HRT with the risk of ovarian cancer. Egger's test also provided no evidence of significant small-study effect for the association of menopausal HRT with the risk of ovarian cancer by histologic subtype .Visual inspection of the funnel plot  and EggeThe current meta-analysis assessed associations of menopausal HRT with the risk of ovarian cancer in various histologic subtypes. Findings of this meta-analysis indicated a positive association of menopausal HRT with the risk of ovarian cancer. In subgroup analysis by study design, significant positive associations were observed in both cohort and case control studies. In subgroup analysis by histologic subtypes, we found that menopausal HRT may increase the risk of serous and endometrioid tumors. In subgroup analysis by the hormones types, significant positive associations were observed for both HRT and ERT. The pooled RR indicated that there might be a stronger association in ERT users, but the result might be not true enough with the insufficient studies about ERT.The mechanism underlying the association of menopausal HRT with ovarian cancer is not well-understood. A theory suggests that high levels of gonadotropins during menopause act as a promoter on the affected ovarian tissue . These fBetween-study heterogeneity is common in meta-analysis. It is necessary to explore the potential sources of between-study heterogeneity. In this meta-analysis, a moderate between-study heterogeneity was found. However, meta-regression analysis with the covariates of study design, published year and continent revealed no source of between-study heterogeneity. After excluding two study   in ovarian cancer analysis, between-study heterogeneity was slightly reduced, and the results did not change substantially, suggesting that they were stable. Fifthly, in subgroup analysis stratified by the hormones types, we found both HRT and ERT could increase the risk of ovarian cancer.However, there were still some deficiencies in this meta-analysis. First, the authors adjusted for confounders such as age, parity, duration of oral contraceptive use, tubal ligation, age at natural menopause and age at menarche etc. in original studies, but we dare not deny whether some unknown confounders might lead to exaggerating or underestimating the association. In addition, confounders adjusted for in various studies were different, which might affect the observed association. Some common biases such as selection bias, recall bias and lost to follow-up etc. in observational studies might also affect the authenticity of the results. Secondly, follow-up durations in various cohort studies differed. Some potential cases might not be observed due to limited follow-up in certain studies. Thirdly, menopausal HRT might be slightly different in each of the papers analyzed. In some papers, the HRT referred to estorgens + progestins, but in others, the HRT referred to only estorgens or estorgens + progestins. This might affect the observed association. Fourthly, the limited amount of studies assessing histologic subtypes made it difficult to confirm the relationship in terms of the kind of therapy and its association with the histological subtypes of ovarian cancer. Fifthly, although the age of subjects was over 50 years old in most included studies, a few studies covered the data from women from pre-menopausal age. This might bias the results of the meta-analysis. Sixthly, the insufficient available data prevented us from conducting a dose-response relationship to explore the association between length of HRT use and the risk of ovarian cancer.In conclusion, this meta-analysis suggests that menopausal HRT may increase the risk of ovarian cancer, especially for serous and endometrioid tumors. This finding requires confirmation by further studies of associations of menopausal HRT with the risk of ovarian cancer in various histological subtypes.All datasets for this study are included in the article/supplementary material.YL and LM conceived and coordinated the study, designed, performed and analyzed the experiments, and wrote the paper. XY, JB, DL, CS, and JZ carried out the data collection, data analysis, and revised the paper. YM and JL designed the study, carried out the data analysis, and revised the paper. All authors reviewed the results and approved the final version of the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Long noncoding RNAs (lncRNAs) have been reported to serve as diagnostic and prognostic biomarkers of cancers, which play vital roles in tumorigenesis and tumor progression. Several studies have been performed to explore diagnostic value of lncRNA H19 in cancer detection and diagnosis. However, there are still inconsistent results in diagnostic accuracy and reliability in individual studies. Therefore, the present study was performed to summarize the overall diagnostic performance of lncRNA H19 in cancer detection and diagnosis. A total of eight studies with 770 cases and 815 controls were included in this pooled analysis. The pooled diagnostic results were as follows: sensitivity, 0.69 (95%CI=0.62-0.76), specificity, 0.79 (95% CI=0.70-0.86), positive likelihood ratio (PLR), 3.31 (95%CI=2.29-4.78), negative likelihood (NLR), 0.39 (95%CI=0.31-0.49), diagnostic odds ratio (DOR), 8.53 (95%CI=4.99-14.60), and area under the curve (AUC), 0.79 (95%CI=0.76-0.83). Deeks' funnel plot asymmetry test (P=0.13) suggested no potential publication bias. Our results indicated that lncRNA H19 had a relatively moderate accuracy in cancer detection and diagnosis. Further comprehensive prospective studies with large sample sizes are urgently required to validate our findings. With incremental incidence and mortality in recent years, cancer has been a major public health problem all over the world , 2. AlthLong noncoding RNAs (LncRNAs) are a subclass of regulatory ncRNAs longer than 200 nucleotides, lacking functional open reading frames (ORFs) and protein-coding capability , 7. LncRH19, a subclass of long noncoding RNA, is a paternally imprinted gene which locates in chromosome 11p15.5 . In receLiterature research was performed in database including PubMed, Web of Science, Wanfang library, and CNKI up to May 18, 2018, by the following searching strategy: “cancer” or “tumor” or “carcinoma” or “neoplasm” or “malignancy” or “neoplasm” and “H19” and “sensitivity” or “specificity” or “ROC curve” or “accuracy”. Three investigators  checked the titles and abstracts of the studies and scanned the full texts to eliminate irrelevant studies with the following included criteria: (1) the diagnostic value of lncRNA H19 for detecting cancer evaluated in articles, (2) explicitly defined article population and control sources; (3) completed data for calculating sensitivity and specificity; and (4) being published in English or Chinese.For each study, the following information was extracted: first author, year of publication, country, ethnicity, sample size, specimen and cancer type, detection method, cutoff value, true positive (TP), false positive (FP), true negative (TN), and false negative (FN).The QUADAS-2 was applied to systematically evaluate the quality of the studies included in this pooled analysis. With the max QUADAS-2 score of 7, we can judge the quality of the included studies based on the results.2) test [2 value higher than 50% indicated obvious heterogeneity between the studies [All statistical analyses were performed using Stata 14.0 . The pooled sensitivity, specificity, diagnostic odds ratio (DOR), positive likelihood ratio (PLR) and negative likelihood ratio (NLR) and other parameters were calculated by the bivariate meta-analysis model. Then, we performed summary receiver operator characteristic (SROC) curves analysis and calculated the area under the ROC curves (AUC) to assess the overall diagnostic value of lncRNA H19 in cancer detection and diagnosis . These d2) test . A P val studies . Moreove studies .By searching PubMed, Web of Science, Wanfang, and CNKI databases, a total of 8 eligible studies –26 incluThe results of the Quality Assessment of Diagnostic Accuracy Studies-2 (QUADAS-2) study quality assessment were also shown in 2=76.41%, 95%CI=60.11%-92.70%) and specificity . In addition, the pooled PLR was 3.31 (95%CI=2.29-4.78), the NLR was 0.39 (95%CI=0.31-0.49), and the DOR was 8.53 (95%CI=4.99-14.60) , and the P value was 0.314 which indicated that the HSROC was symmetrical. The value of γ was 2.08 (95%CI=1.56-2.60)  . To evalGod-of-fit and bivariate normality analyses Figures  showed t2 of the heterogeneity test was 89%, indicating significant heterogeneity. In the present study, the calculated Spearman correlation coefficient value was −0.11 with p=0.01 (P<0.05), suggesting that the threshold effect was the major source of heterogeneity. However, there were only eight articles included; metaregression analysis and subgroup analysis cannot be used.The IThrough the next generation sequencing technology and large-scale transcriptome mapping, many lncRNAs have been reported to involve in the development of cancer as a regulator in a variety of biological processes. These lncRNAs, located in the nucleus, interact with chromatin remodeling complexes (CRCs) to regulate the genes expression locating on the same chromosome in cis or on another chromosome in trans through fine-tuning of chromatin architecture , 35. PreMuch effort has been made to investigate the link between aberrant lncRNA expression and cancer, including lncRNA H19 , 39–41. Heterogeneity is an inescapable problem that can interpret the results of the meta-analysis . There wNevertheless, several defects of this pooled analysis should be emphasized. First, eight studies with a limited number of subjects were included in this study, which may weaken the reliability for determining the diagnostic value of H19 for different types of cancers. Second, our articles have a very high ratio of data in Chinese populations, which may result in inevitable publication bias. Third, research and sample size in single tumor type was relatively small; more cancer types studies with large sample size need to be included in analysis. Fourth, not all of the studies reported the cutoff values of lncRNA H19. Finally, only publications in English or Chinese were included; researches in other languages should not be missed.In summary, all of the results indicated that H19 had a relatively moderate accuracy in distinguishing cancer patients from all individuals, suggesting that H19 could serve as a potential diagnostic biomarker for cancer detection and diagnosis. Furthermore, well-designed prospective studies with large sample sizes and different population groups must be conducted in the future to confirm our findings.


The association between vascular endothelial growth factor (VEGF) gene polymorphisms  and lung cancer risk has been extensively studied in the last decades, but currently available results remain controversial or ambiguous. Therefore, we conducted a meta-analysis to assess whether the relationship between the VEGF gene and lung cancer susceptibility exists. The meta-analysis was conducted by searching the databases PubMed, Embase, and Web of Science covering all eligible studies published up to October 1, 2017. The pooled odds ratios (ORs) as well as their 95% confidence intervals (CIs) were utilized to evaluate the possible associations. Publication bias of relevant studies was examined via Begg's funnel plots and Egger's regression tests. P = 0.184; homozygote model: pooled OR = 1.51, 95% CI: 1.12–2.03, P = 0.821), but no significant results were detected in Caucasian populations. This meta-analysis included 13 published case–control studies covering 4477 patients with lung cancer and 4346 healthy controls, who had been accrued from December 1992 to July 2012. For the overall eligible data collected in our meta-analysis, it indicated that VEGF +936C/T, -460C/T, and -2578C/A polymorphisms did not correlate with the elevated lung cancer risk in all genetic comparison models. Moreover, VEGF +460T/C polymorphism was found to be significantly associated with susceptibility to lung cancer in these models (allele model: pooled OR = 1.12, 95% CI: 1.00–1.26, VEGF +936C/T, -460C/T, and -2578C/A polymorphisms were not associated with the risk of lung cancer. The VEGF +460T/C polymorphism might be a risk factor for lung cancer only in Asian populations. Lung cancer, as one of the most frequently occurring cancers in the world, is always the leading cause of cancer-related death among both men and women . Approxi in vitro and in vivo [Angiogenesis refers to the complex process of the formation of new networks of blood vessels. Increased angiogenesis has been proved to be associated with the process of tumor growth and metastasis . Vascula in vivo , 10. Ser in vivo . Moreove in vivo . In lung in vivo . The VEG in vivo . Targete in vivo , 16.The VEGF gene is located at chromosome 6p21.3, covering 14 kb in length with 8 exons and 7 introns . Being hA systematic literature search was conducted on PubMed, Embase, and Web of Science databases until October 1, 2017, to identify studies for inclusion in the meta-analysis. The relevant key words and search strategies were as follows: “vascular endothelial growth factor” or “VEGF,” “polymorphisms” or “mutation,” “variants” and “lung cancer,” “non-small cell lung cancer,” or “NSCLC.” These terms were arranged into different combinations when used for the search. Besides, the reference lists of original studies were searched manually for additional literature. All the eligible studies were checked carefully to prevent overlapping datasets, and only previously published studies were included.The studies that fulfilled the following criteria were considered eligible: (1) studies focused on the association between VEGF and the risk of lung cancer; (2) studies used a case–control or cohort design; and (3) inclusion of sufficient data on the frequency of genotypes including ORs and their 95% CIs. In addition, the studies that did not meet the aforementioned inclusion criteria were excluded.The data were independently extracted from all the eligible studies by two investigators (Yang FM and Qin ZQ) according to the aforementioned inclusion criteria. When it came to conflicting evaluations, an agreement was settled after discussion with the third reviewer. For each study, the extracted information included: the surname of the first author, year of publication, ethnicity, source of controls, sample size of case and control groups, SNP genotyping methods, genotype distributions, and the results of the Hardy–Weinberg equilibrium (HWE) test. P value <0.05 was considered to indicate statistically significant heterogeneity.The pooled odds ratios (ORs) and their corresponding 95% CIs were calculated to evaluate the strength of association between VEGF gene polymorphisms and lung cancer risk under five genetic models: allele model, homozygous model, heterozygous model, dominant model, and recessive model. The genotypic distributions of different polymorphic sites in the subjects were compared with those expected under HWE using the chi-squared goodness-of-fit test, and aP < 0.05) or absence (P > 0.05) of heterogeneity. When the P value was >0.05, the pooled ORs were calculated using the fixed-effects model based on the Mantel–Haenszel method. Otherwise, the random-effects model with the DerSimonian–Laird method was used for this meta-analysis. Then, the sources of heterogeneity were further analyzed according to ethnicity. The sensitivity analysis was performed by sequentially excluding individual studies to evaluate the stability and reliability of the results. In addition, publication bias between the studies was analyzed using Begg's funnel plots and Egger's linear regression tests. All the aforementioned statistical tests were performed using Stata12.0 software.The pooled ORs were calculated using the fixed-effects model or the random-effects model according to the presence  for the allele model, 1.32  for the homozygote model, 1.11  for the heterozygote model, 1.12  for the dominant model, and 1.25  for the recessive model, indicating no association between VEGF +936C/T mutation and lung cancer susceptibility , homozygote model , heterozygote model , dominant model , and recessive model   . When th= 0.151) . P < 0.001) for the allele model, 1.15  for the homozygote model, 1.33  for the heterozygote model, 1.33  for the dominant model, and 0.98  for the recessive model, indicating no significant association between VEGF -2578C/A polymorphism and lung cancer risk . Result from Begg's test and Egger's test also confirmed the absence of publication bias. Considering the above tests, the results of our meta-analysis were reliable.The possible publication bias of the studies involved in the present meta-analysis was examined via Begg's funnel plot and Egger's test. As shown in As an important regulator of various malignancy angiogenesis, VEGF is always involved in the development and progression of multiple cancers . The proThis meta-analysis included 13 independent case–control studies with a total of 4477 patients with lung cancer and 4346 healthy controls to investigate the correlation between VEGF variants and the risk of lung cancer. The results of our meta-analysis showed that VEGF +936C/T, -460C/T, and -2578C/A gene polymorphisms were not associated with the risk of lung cancer. However, the conclusion was still inaccurate due to limited published data. Furthermore, the predictive value of VEGF polymorphisms in the prognosis of lung cancer could not be assessed. The sample size in the present meta-analysis was larger than that in any individual study, leading to more precise and robust results. Thus, the present meta-analysis aimed to provide a more powerful and reliable conclusion about the relationship between VEGF genetic polymorphisms and lung cancer risk. Moreover, no statistically significant association was detected between VEGF +936C/T, -460C/T, and -2578C/A polymorphisms and lung cancer risk in all genetic comparison models. Interestingly, the VEGF +460T/C polymorphism was found to be significantly associated with the susceptibility to lung cancer only in Asian populations. Although the exact mechanism underlying ethnic differences was unknown, the possible reasons could be the natural selection pressures or a balance by other related functional genetic variants and environmental exposures, resulting in gene polymorphisms [However, the present meta-analysis had several limitations. First, the number of included studies in our meta-analysis was not so abundant, leading to limited statistical precision, especially in subgroup analysis. Hence, more studies using standardized unbiased methods are needed to offer more detailed individual data. Second, the population included in this meta-analysis is mainly limited to Asian and Caucasian ethnicities, and more data from other ethnic groups will be required. Third, since only data from previously published studies were included in the meta-analysis, it was possible that some related published studies or unpublished studies had null results, resulting in a deviation from the expected results. Finally, tumorigenesis is a complex process involved in the regulation of a series of genetic factors besides VEGF. As a multifactorial disease, the risk of developing lung cancer was closely related to the environment, smoking, occupational exposure, and interaction among various genetic factors, and not just a single factor. Therefore, more raw data are needed to adjust other variables, such as age, sex, and smoking. Consequently, further high-quality studies on the risk of lung cancer should be conducted over the next few years to achieve more accurate results. gene polymorphisms and lung cancer risk.In sum, this meta-analysis found that VEGF +936C/T, -460C/T, and -2578C/A polymorphisms were not associated with the susceptibility to lung cancer. Interestingly, the VEGF +460T/C polymorphism might be a risk factor for lung cancer only in Asian populations. Based on our discovery, additional large population-based multicenter prospective studies are needed to confirm the association of VEGF


The relationship between male pattern baldness and incidence of testicular cancer remains inconclusive. Hence, we performed the present meta-analysis based on all eligible case–control studies.A comprehensive literature search was performed in July 30th 2018 based on PUBMED, EMBASE and Web of science database. Pooled odds ratio(OR) and its 95% confidence intervals (95% CIs) was calculated with a DerSimonian and Laird random-effects.P = 0.22 for heterogeneity, I2 = 30%). When subgroup analysis by types of baldness, We found a statistically significant association was observed that baldness(I-VII) might become a protective factor for the risk of testicular germ cell tumor(TGCT). There was no definite connection between alopecia and the different types of TGCT.The pooled results were included in this meta-analysis. Overall, We have demonstrated statistically signification between baldness(any pattern) and testicular cancer was identified . There was no obvious heterogeneity across included studies  hair loss can decrease more strongly testicular cancer risk than any other grades. Despite of our findings, We still need further researches to advance knowledge in this field. Male pattern baldness, also known as Androgenetic alopecia (AGA), is an inherited pattern of baldness hormone-regulated by the testosterone metabolite dihydrotestosterone characterized by follicular androgen sensitivity and follicular miniaturization . While iAs a result, we conducted a systematic review and a comprehensive meta-analysis in order to further investigate the issue and identify potential sources of heterogeneity that might be confounders that have affected some existing conclusions.A comprehensive literature search was performed in JULY 30th 2018 based on PubMed, EMBASE and Web of Science databases with the following search algorithm:  and . In addition, the reference lists of retrieved articles and related reviews were checked to identify any potential relevant studies. No language or publication date restrictions were adopted. The flow diagram was presented in Fig. Articles included in this study had to meet all the following criteria: the exposure of interest was male pattern baldness; the outcome of interest was risk of testicular cancer; study design was nested case–control, or case-control; and the risk estimates with their corresponding 95% CIs were reported or sufficient datas were provided to calculate them.Newcastle–Ottawa Scale (NOS) was introduced to evaluate the quality of included studied by two independent reviewers. NOS is a 9-star system, which includes 3 dimensions: selection (4 items), comparability (1 item), and exposure/outcome (3 items). Each item represents 1 point, except for comparability (2 points). A study with ≥7 points was considered as a high-quality study.Data were extracted independently by the same 2 reviewers  and any discrepancies were resolved by discussion. The following data from each included study were recorded: first author’s surname, the country where the study was conducted, study design, publication year, sample size and number of cases, types of baldness, fully adjusted risk estimates with their corresponding 95% confidence intervals (95% CIs), and matched or adjusted variables in the study design or statistical analysis.2 score was also used to determine the degree of heterogeneity [For studies that reported separate effect size estimates for different types of baldness, we combined these risk estimates within each study, weighted by inverse of the variance . Subgrougeneity) . Galbraigeneity)  was intrIn sensitivity analysis, meta-analysis was reconducted after omitting each study in turn. Potential publication bias was evaluated with Begg’s test and Egger’s test . Statistn = 2), Italy (n = 1), Greece (n = 1), Nation (n = 1), There were 5 case–control studies [The detailed process of literature search is shown in Fig.  studies , 14–17, P < 0.00001)    Fig. . When th67) Fig. . When th01) Fig. .Fig. 2FoSensitivity analysis confirmed that no individual study influenced the overall results. There was no evidence of publication bias in this meta-analysis indicated by the Begg’s funnel plot and Egger’s tests  Fig. 4)4).Fig. 4p = 0.04). This was the first time that baldness had been linked to testicular cancer in a case-control study. However,in this study report,it had some limitations, including that it only considered the relationship between different types of alopecia and testicular cancer,but not included the subtypes of testicular cancer were not presented. In Farzana et al. [This systematic review and meta-analysis involved a total of 884 cases from 5 observational studies , 14–17. a et al. , they foa et al. , they goa et al. , in addia et al. , accordia et al.  they gota et al. . It has a et al. , It has a et al. . In addia et al. . These fIn the similar study reported by Liang et al. , we got As an important organ that maintains male charecteristics, the human testes are prime targets for hormonal modulation. We found evidence of association between AGA at any age and a significant decreased risk of testicular cancer based on meta-analysis of case-control studies,in histological subgroup analysis, we explored that the second level of hair loss is more benificial for reducing the risk of TGCT. This conclusion may suggest that AGA or men with baldness may be the protective factor for the TGCT. AS far as we concerned, our study includes only 5 case-control reports and there are some divergent views in these documents. Further andomized controlled trials studies should be required to confirm the potential relationship between AGA and TGCT.


The pooled ORs  of GST M1 and T1 were 1.186 and 1.107, respectively. In addition, subgroup analyses according to the type of SNHL and ethnicity showed no relationship between GST M1 and T1 polymorphisms and the susceptibility to acquired SNHL. Our results suggest that no significant relationship was found between GST M1 and T1 polymorphisms and the susceptibility to acquired SNHL.Acquired sensorineural hearing loss (SNHL), including age-related hearing loss (ARHL), noise-induced hearing loss (NIHL), drug-induced hearing loss (DIHL) and sudden sensorineural hearing loss (SSHL), is one of the most common sensory deficits in humans. Several studies have reported that antioxidant gene glutathione s-transferase M1 and T1 (GST M1 and T1) polymorphisms have a close relationship with the susceptibility to acquired SNHL, but other articles have reported opposite results. This meta-analysis aims to identify whether an association exists between GST M1 and T1 polymorphisms and the susceptibility to acquired SNHL. Seventeen independent studies containing 1749 cases and 2018 controls were included. According to the I Approximately 360 million people worldwide suffer from this health problem2. People with acquired SNHL exhibit decreased hearing sensitivity and a decline in speech intelligibility, which can lead to serious difficulties in an individual’s communication and social interactions and consequently reduce life expectancy4. Despite the high prevalence and serious effects of acquired SNHL, few therapeutic methods have been found to be clinically effective5.Acquired sensorineural hearing loss (SNHL), including age-related hearing loss (ARHL), noise-induced hearing loss (NIHL), drug-induced hearing loss (DIHL) and sudden sensorineural hearing loss (SSHL), is one of the most common sensory deficits in humans in modern society10. Therefore, exploring these genetic differences and then exploiting them may enable the development of individual prevention strategies for SNHL.The susceptibility to acquired SNHL among individuals is diverse. Some individuals are more susceptible to acquired SNHL, while others are not. Several studies have suggested that this individual difference in susceptibility to acquired SNHL is mostly due to the different genetic backgrounds of individuals, especially genetic polymorphisms that affect the expression of some functional proteins or enzymes14. Researchers have successfully alleviated several kinds of SNHL with the application of antioxidants in animal experiments15. Glutathione s-transferase (GST) encodes a system of antioxidative enzymes that have been demonstrated to play an important role in antioxidative protection in cochlear cells18. Among the GST subclasses, GST T1 and M1 are genetically deleted (null genotype) in a high percentage in humans. Approximately 30–50% of individuals have a null genotype for GST M1, depending on their race19, and 25–40% carry the null genotype of GST T120. Rabinnowitz et al. once suggested that individuals with the null genotypes of GST M1 or GST T1 are more susceptible to oxidative stress damage and are possibly more susceptible to NIHL21.Oxidative stress has been proven to be the most important molecular mechanism in the pathogenesis of acquired SNHL25, and others have reported conflicting results34. Meta-analysis is an effective way to address this type of contradiction. Therefore, we performed this meta-analysis to identify whether a close association exists between GST M1 and T1 polymorphisms and the susceptibility to acquired SNHL and whether GST M1 and T1 polymorphisms can serve as predictive factors for the susceptibility to acquired SNHL.Many studies have attempted to correlate mutant genotypes of GST to the susceptibility to acquired SNHL. Some have demonstrated a close relationship between GST M1 or T1 polymorphisms and the susceptibility to SNHLThe literature selection process is shown in Fig. 2 values for GST M1 and T1 were 50.9% and 64.3%, respectively, which are shown in Fig. 2 values were ≥30%, so we used a random-effects model to calculate the pooled odds ratios (ORs) and 95% confidence intervals (95% CIs). The pooled ORs  of GST M1 and T1 were 1.186 and 1.107, respectively.The ITo address heterogeneity, we performed a subgroup analysis according to the acquired SNHL type and ethnicity. For GST M1, no heterogeneity was observed in the DIHL and the SSHL subgroups. However, relatively strong heterogeneity was observed in the ARHL and NIHL subgroups  most of the previous studies were single-center studies with small sample sizes; and (2) most of the previous studies focused on the associations of GST M1 and T1 polymorphisms with only one type of acquired SNHL and neglected the fact that oxidative stress is the most important and common molecular mechanism of acquired SNHL. Based on the above theoretical basis, this meta-analysis included more studies on oxidative stress-related acquired SNHL types and larger sample sizes to further identify the association of GST M1 and T1 polymorphisms with the susceptibility to acquired SNHL. According to our results, neither the collective nor the subgroup analyses suggested an association between GST M1 and T1 polymorphisms and susceptibility to acquired SNHL.Oxidative stress is the most important and common molecular mechanism of acquired SNHLet al. in 2014. The study concluded that GSTM1polymorphisms, but not GST T1 polymorphisms, are related to noise-induced hearing loss. There are at least two major differences between Zhou’s study and our study. (1) Zhou et al. evaluated the association of GST M1 and T1 polymorphisms only with NIHL. Five studies were included in their meta-analysis, with a total of (GST M1 and T1) 914/909 cases with NIHL and 885/876 controls. However, the purpose of our study was to identify the possible relationship between GST M1 and T1 polymorphisms and acquired SNHL, which includes three other kinds of SNHL in addition to NIHL, with a total of 1749 cases and 2018 controls included in our meta-analysis. (2) Zhou et al. concluded that GST M1 polymorphisms are related to NIHL, but in our study, we found opposite results, although the included studies and samples in the NIHL subgroup in our study were the same as those included in their study. The different results can be attributed to the different models used in the two studies. In Zhou’s study, the fixed-effects model was applied to calculate the pooled effect size even though significant heterogeneity (51%) was observed among the studies. Such a method is worth discussing.A meta-analysis regarding the association between GST M1 and T1 polymorphisms and NIHL was performed by Zhou Heterogeneity is a major problem that affects the reliability of the pooled effect size in meta-analysis. In our meta-analysis, heterogeneity was observed for both GSTM1 and GST T1. The results of the subgroup analysis according to the type of acquired SNHL and ethnicity showed that heterogeneity was much smaller in some subgroups, but it was strong in other groups, suggesting that some other factors besides the acquired SNHL type and ethnicity served as sources of heterogeneity. We also performed a meta-regression analysis  Several studies that possibly met the inclusion criteria did not include primary data, so the ORs with their 95% CIs could not be calculated. We attempted to contact the authors for more information, but we received no response. The results may be influenced by these missing studies. (2) Although 17 independent studies containing a total of 1749 cases with acquired SNHL and 2018 controls were included in this meta-analysis, the sample size is still limited, especially in the process of subgroup analysis. For the DIHL subgroup, most studies contained only 10–20 samples, and for the SSHL subgroup, only 2 articles met the inclusion criteria.To our knowledge, this is the first meta-analysis to focus on the association of GST M1 and T1 polymorphisms with the susceptibility to acquired SNHL. The results of our meta-analysis suggested that GST M1 and T1 polymorphisms may not serve as susceptibility factors for acquired SNHL. Considering the limitations of our meta-analysis, further prospective studies with large sample size and additional studies  are needed to validate study findings.A comprehensive literature search was performed in the following databases: (1) PubMed; (2) Web of Science; (3) EMBASE; (4) OVID; (5) CNKI Chinese database and (6) Wanfang Chinese database. The MeSH and free terms were all included in our search terms, which are listed as follows: “Glutathione s-transferase”, “Glutathione transferase”, “hearing impairment”, “hearing loss”, “ototoxicity” and “deafness”. Our search logic in the PubMed database is listed as follows: “((((“hearing”[MeSH Terms] OR “hearing”[All Fields]) OR  OR “inner ear”[All Fields] OR “cochlea”[All Fields] OR “cochlea”[MeSH Terms])) OR ototoxicity[All Fields]) OR (“audiology”[MeSH Terms] OR “audiology”[All Fields])) AND (((“glutathione transferase”[MeSH Terms] OR (“glutathione”[All Fields] AND “transferase”[All Fields]) OR “glutathione transferase”[All Fields] OR “glutathione s transferase”[All Fields]) OR (“Glutathione Transferase”[Mesh] AND “Glutathione S-Transferase pi”[Mesh] AND “glutathione S-transferase T1”[Supplementary Concept] AND “glutathione S-transferase M1”[Supplementary Concept])) OR (“glutathione transferase”[MeSH Terms] OR (“glutathione”[All Fields] AND “transferase”[All Fields]) OR “glutathione transferase”[All Fields])) AND “humans”[MeSH Terms]”.th, 2018. We also manually checked all articles listed in the reference lists of the retrieved literature.All studies that we searched were published before November 20Studies that met the following criteria were included: (1) independent studies investigating the relationship between GST M1 and T1 polymorphisms and the susceptibility to acquired SNHL and (2) studies including sufficient and definite original data (the genotype frequencies of GST M1 and T1 in the case and control groups) that could be used to calculate the OR with its 95% CI of each genotype. When duplicate publications were found, the data in the latest publication were used.The data in the included studies were extracted by two investigators independently using the same “Data Extraction Form”. The information extracted from the included studies is listed as follows: first author’s name, publication year, country of origin, ethnicity, genotype detection methods, the type of SNHL, and the number of cases and controls. The quality of each included study was evaluated using the Newcastle-Ottawa scale (NOS). The studies with an NOS score ≥7 were considered high-quality studies. All disagreements in the process of study selection, data extraction and quality assessment were discussed and resolved by consensus.2 test. For a value of I2 < 30% and p > 0.1, a fixed-effects model was used to calculate the pooled ORs; otherwise, a random-effects model was used for a value of I2 ≥ 30%.Woolf’s method was applied to estimate the 95% CIs. We considered that there was statistical significance when the overall 95% CI did not include 1 and the p-value transformed from the Z score was less than 0.05. In addition, a subgroup analysis was performed according to the type of acquired SNHL and ethnicity. Sensitivity analysis was used to evaluate the stability of the pooled effect size. Publication bias was assessed by Egger’s test. Publication bias was considered for a p-value < 0.05 or if the 95% CI did not contain 0. All statistical analyses were performed using the Stata 13.1 software.The association of GST M1 and T1 polymorphisms with acquired SNHL susceptibility was evaluated by the pooled OR and 95% CI.Statistical heterogeneity among the studies was measured with the IResults of meta-regression considering sample size, publication year and NOS scores of the included studies.