MIR499A rs3746444 polymorphism with breast cancer susceptibility, but the results have been inconsistent. In this work, we performed a meta-analysis to obtain a more reliable estimate of the association between the polymorphism and susceptibility to breast cancer. A comprehensive literature search was conducted on PubMed, Scopus, Web of Science (WoS), China National Knowledge Infrastructure (CNKI), VIP and Wanfang databases up to January 2020. A total of 14 studies involving 6,797 cases and 8,534 controls were included for analysis under five genetic models: homozygous (GG vs. AA), heterozygous (AG vs. AA), dominant (AG + GG vs. AA), recessive (GG vs. AA + AG) and allele (G vs. A). A statistically significant association was observed between the polymorphism and an increased breast cancer susceptibility under all genetic models . Subgroup analysis based on ethnicity suggested that significant association was present only among Asians, but not Caucasians. In conclusion, MIR499A rs3746444 polymorphism was significantly associated with breast cancer susceptibility among Asians, suggesting its potential use as a genetic risk marker in this population.Numerous studies have investigated the association of  Early identification of individuals at risk of the cancer is the key to its prevention. Currently, genetic testing has emerged as a promising strategy for predicting breast cancer risk. The most regularly tested genes in breast cancer are BRCA1 and BRCA2, whose mutations have been unequivocally linked to a substantially elevated risk of the cancer. However, the overall prevalence of these high penetrance mutations is low, ranging from 0.2–0.3% in the general population to approximately 20% in breast cancer patients with a family history of the malignancy2. Thus, mutations in BRCA genes have a relatively low population attributable risk (PAR) and as such, the benefit of screening for these mutations at a population-wide scale is largely debatable3.Breast cancer is the most common type of cancer and the leading contributor to cancer-related deaths in women worldwide4. In contrast to high penetrance mutations, genetic polymorphisms are very common in the general population, but are typically associated with a modest risk of cancer. However, it is believed that when a large number of such polymorphisms are combined, the magnitude of the risk conferred could be very significant5. In recent years, polymorphisms in microRNA genes have been widely investigated as the gene products play important roles in regulating the expression of many cancer-related genes8. Among the microRNA genes that have been frequently investigated is MIR499A, which encodes microRNA-499a. The functions and mechanism of miR-499a in breast cancer is not well understood, but it is known that miR-499a possesses both oncogenic and tumor suppressive functions in other cancer types. For example, in colorectal cancer, miR-499a could promote invasiveness and metastatic capabilities by targeting FOXO4 and PDCD4 tumor suppressor genes9. Similarly, in hepatocellular carcinoma (HCC), the microRNA was found to target MAPK6 to induce proliferation and migration of the cells11. However, in HCC, the microRNA was also observed to target the ETS1 oncogene to inhibit the cancer progression12. Additionally, in non-small cell lung cancer, miR-499a was found to exert its tumor suppressive function by targeting VAV3 oncogene13.Over the past decades, the potential of genetic polymorphisms as markers for breast cancer risk assessment has become increasingly apparent14. Thus, polymorphisms occurring in the seed region of a microRNA gene may contribute to its oncogenic or tumor suppressive functions and subsequently affect cancer risk or susceptibility. One such polymorphism is rs3746444, which results from an A-to-G substitution in the seed region of MIR499A. A number of studies have examined the association of the polymorphism with breast cancer susceptibility, but the results obtained have been inconsistent. For instance, while He et al.15 demonstrated that the variant allele of MIR499A rs3746444 polymorphism was significantly associated with an increased susceptibility to breast cancer, Morales et al.16 did not find any significant association. One of the major reasons for the inconsistency of study results in genetic association studies is the limited sample size and thus, low statistical power of the individual studies18. Therefore, in this work, we aimed to pool the study findings via a meta-analysis in order to derive a more precise estimate of the association between MIR499A rs3746444 polymorphism and breast cancer susceptibility.The mechanism of microRNA target recognition and selection is determined, at least in part, by its seed sequences, i.e. a conserved region essential for its binding to target mRNAsA comprehensive literature search was performed on PubMed, Scopus, Web of Science (WoS), China National Knowledge Infrastructure (CNKI), VIP and Wanfang databases up to January 2020. No language restriction was imposed. The PubMed search terms used were as follows: [(breast neoplasms OR breast cancer) AND (MIR499A OR MIR-499 OR MIR-499A OR microRNA-499 OR microRNA-499A OR rs3746444) AND (polymorphism OR mutation OR variant OR variation)]. The search terms were modified appropriately for other databases.MIR499A rs3746444 polymorphism and breast cancer susceptibility; (ii) case-control (including nested case-control) or cohort in design; and (iii) contained data on genotype and allele frequencies or sufficient data to derive this information. Non-human studies and duplicated reports were excluded. When overlapping data were found in multiple studies, only the most recent report or the one with the largest sample size was included.Eligible studies were selected based on the following criteria: (i) those that investigated the association between 19. Studies with ≥5 stars were considered to be of high quality.Two investigators independently extracted the following information from eligible studies: First author, year of publication, country, number of cases and controls, genotype and allele frequencies, ethnicity of the subjects (Asian vs. Caucasian), and deviation of control genotype distribution from the Hardy-Weinberg equilibrium (HWE). If data on HWE was not reported, a goodness-of-fit test was used to determine whether the observed genotype frequencies conformed to the expected distribution. The quality of the eligible studies was assessed by using the Modified Newcastle-Ottawa Scale for Case-Control Studies of Genetic AssociationMIR499A rs3746444 polymorphism and breast cancer susceptibility was evaluated under five genetic models: homozygous (GG vs. AA), heterozygous (AG vs. AA), dominant (AG + GG vs. AA), recessive (GG vs. AA + AG) and allele (G vs. A) comparison models. Heterogeneity among the studies in each genetic model was assessed by using Cochran’s Q and I2 statistics. A P value of <0.1 and I2 value of >50% was interpreted as having significant heterogeneity. A fixed-effect method was used to calculate the pooled odds ratio (OR) and the corresponding 95% confidence interval (CI) for genetic models which did not show significant heterogeneity; otherwise a random-effect method was used. The significance of the genetic association was measured using Z test. A forest plot was constructed to graphically display the results. A subgroup analysis was performed based on ethnicity of the subjects (Asian vs. Caucasian) and methodology quality of the studies . Sensitivity analysis was performed by iteratively omitting one study at a time to determine the stability and robustness of the results. Publication bias was assessed by performing Begg’s test and Egger’s test using JASP (version 0.9.2.0), and by visually inspecting the funnel plot for asymmetry. For all analyses, statistically significance was assumed at P < 0.05, unless otherwise stated.Quantitative synthesis of the data was performed by using Review Manager (version: 5.3.5), with the common allele/genotype used as the reference. The association between http://diana.imis.athena-innovation.gr/DianaTools/index.php?r=tarbase/index) and DIANA microT-CDS (http://diana.imis.athena-innovation.gr/DianaTools/index.php?r=microT_CDS/index) tools which have been integrated with miRBase (ver. 18) and Ensembl (ver. 69). Functional annotation of the predicted target genes as well as KEGG pathway enrichment analysis was then performed by using DIANA-miRPath version 3.0 (http://snf-515788.vm.okeanos.grnet.gr/)23. The DIANA-microT threshold used was 0.8 at a P value of 0.05. Fisher’s exact test (hypergeometric statistics) and false discovery rate (FDR) correction was used for the enrichment analysis. miRmut2Go (http://compbio.uthsc.edu/miR2GO/mir2goSNP.php) was used to compare the functional similarity between the wild type and variant alleles of the MIR499A rs3746444 polymorphism. Furthermore, RNAfold WebServer (http://rna.tbi.univie.ac.at/cgi-bin/RNAWebSuite/RNAfold.cgi) was used to predict, based on minimum free energy (MFE) calculation, the secondary structures formed as a result of the A-to-G substitution. PolymiRTS Database 3.0 (http://compbio.uthsc.edu/miRSNP/) was used to predict whether the polymorphism affects its target recognition.Prediction of miR-499a target genes was performed using DIANA Tarbase version 7.0 . After deduplication, 932 unique records were screened by title and abstract for relevance. A total of 15 records were identified as being potentially relevant; thus, full-text articles of 16 potentially relevant records were retrieved and checked for eligibility. The reference list of these records were also hand-searched to identify additional studies. Ultimately, 12 records (which comprised 13 studies) were included in the meta-analysisThese 14 studies involved a total of 6,797 cases and 8,534 controls. Detailed characteristics of the included studies are shown in Table MIR499A rs3746444 polymorphism with breast cancer susceptibility is summarized in Table The pooled association of et al.30 was omitted. A similar observation was documented in homozygous and allele models when data from Hu et al.24 was removed. In addition, the homozygous model was also heavily influenced by Qi et al.29, while the allele model was heavily influenced by Doulah et al.31, He et al.15, Omrani et al.28 and Qi et al.29 Despite this, the direction and magnitude of the association were not affected by any of the studies, and the P values were still at the borderline of statistical significance.For heterozygous and recessive models, the results remained largely unchanged when any of the studies was removed from the analysis. However, in other models, the results were less stable. In particular, the statistical significance of homozygous, dominant and allele models diminished when data from Dai Funnel plots for publication bias are shown in Fig. The DIANA Tarbase v7.0 and DIANA microT-CDS identified a total of 2,286 target genes for miR-499a. Of these, 1,169 were targeted by the 3p arm, 899 by the 5p arm, and 218 by both arms of miR-499a. Pathway enrichment analysis of these target genes revealed that transcriptional misregulation in cancer is the most common pathway involved (28 genes), followed by RNA transport (26 genes), ubiquitin mediated proteolysis (24 genes), cell cycle and hippo signaling pathway (22 genes)  and molecular functions (similarity score: 0.221). No similarity score could be generated for cellular component as there was no significantly enriched GO term. In addition, analysis with RNAfold WebServer found that the rs3746444 polymorphism did not lead to significant effect on the folding  of the microRNA.MIR499A rs3746444 polymorphism, for example, has been shown to be significantly associated with risk of cancer of the respiratory, digestive, urinary and gynecological systems34. Over the past decade, a growing number of studies have examined the association between MIR499A rs3746444 polymorphism and breast cancer susceptibility, but the results were inconsistent and often contradictory. In this work, we addressed this irregularity by pooling data from 14 previous studies which comprised 15,331 subjects  via a meta-analysis. We found that the polymorphism was significantly associated with an increased breast cancer susceptibility under all five genetic models investigated.MicroRNAs are an emerging class of post-transcriptional regulators which have been implicated in oncogenesis. Polymorphisms within microRNA genes may have an impact on an individual’s susceptibility to cancer. The 26. It has also been reported in an in vitro study that miR-499a originating from the two alleles of rs3746444 polymorphism reduced the expression of tumor suppressor genes to different extents35. Considering these previous findings, we postulate that the variant allele of the polymorphism gave rise to a higher level of miR-499a and caused a significant decrease in the expression of tumor suppressor genes, thereby leading to an increase in breast cancer susceptibility. Besides, our in silico analysis showed that the A-to-G substitution in the rs3746444 polymorphism disrupted the target sites and created new target sites for a large number of genes. Genes targeted by the different alleles of the polymorphism are involved in different biological processes and have different molecular functions. We postulate that the A-to-G substitution resulted in a higher affinity of miR-499a for tumor suppressor genes, which could explain the increased breast cancer susceptibility associated with the variant allele. Nonetheless, the above postulations require further investigations as miR-499a is known to regulate not only tumor suppressor genes, but also oncogenes13.One of the possible explanations for this observation is that the two alleles of the rs3746444 polymorphism resulted in different levels of miR-499a. In fact, it has been demonstrated previously that the variant GG genotype of the polymorphism had the lowest delta CT value (which implied a high expression level), followed by AG and AA genotypes, although it was not clear whether the difference was statistically significant37. Moreover, genetic association is known to be affected by gene-gene, gene-environment and gene-nutrient interactions, which might explain the disagreement in study findings between the two subgroups39. It is also worthy of mention that among the 14 studies included in the meta-analysis, only three were conducted among Caucasians. Hence, it remains a possibility that the lack of significance among Caucasians was a false negative observation due to a relatively weak statistical power40. Only one of the included studies was conducted among Africans; thus, subgroup analysis was not performed on Africans.In the present meta-analysis, a random-effect method was used in several genetic models as significant heterogeneity was observed among the studies. We addressed the heterogeneity by performing subgroup analysis according to the ethnicity of the subjects (Asians vs. Caucasians) as well as the methodological quality of the included studies. In most genetic models, subgroup analysis by ethnicity reduced the statistical heterogeneity. Interestingly, we also observed that significant association was only present in Asians, but not in Caucasians, under most genetic models. This discrepancy could be attributed to the heterogeneity in linkage disequilibrium among subjects of different ethnicities, which is a common concern in genetic association studies involving multiple populations19. It is unknown which of the criteria contributed to the heterogeneity. However, all low quality studies did not fulfill and/or report criteria (v), (vi), (viii) and (ix) – the former two were related to ethnic descent, which again highlighted the influence of ethnicity on the genetic association.Besides, we noted that statistical heterogeneity was eliminated when only studies of high methodological quality were included in the analysis. A high quality study was defined as one which reported and fulfilled at least five of the following criteria: (i) cases were confirmed as having the malignancy, (ii) no evidence of selection bias among cases, (iii) population-based controls and genotypic distribution that conformed to HWE, (iv) appropriate control selection, (v) cases and controls were of the same ethnicity, (vi) no evidence of population stratification, (vii) genotypes validated using an independent method and investigators were blinded to the subject status, (viii) used the same genotyping method for all subjects, and (ix) genotyping call rate >99%41. However, there are inherent limitations associated with each test. Begg’s test does not take into account between-study heterogeneity and is more accurate when the number of included studies is large42. On the other hand, Egger’s test tend to give false-positive results and is more suitable for studies with continuous outcomes44. We included the two tests in our analysis to complement the limitations of each other. Both tests indicated that there was no significant publication bias in all genetic models.In this work, two formal tests were employed to assess publication bias, namely Begg’s and Egger’s tests. Both tests are commonly used in meta-analysis of genetic association studiesMIR499A rs3746444 polymorphism with breast cancer susceptibility among non-Asian populations remained unclear. Nevertheless, a major strength of the present work is that it included a larger number of studies and subjects compared to previous reports on this topic46. Thus, the present meta-analysis provided an updated and integrated estimate of the association between the polymorphism and breast cancer risk. In addition, we performed in silico analysis to predict the functional impact of the polymorphism, which may help to clarify the mechanisms by which MIR499A rs3746444 influences breast cancer susceptibility.There are several limitations of this meta-analysis. First, we reported only crude estimates of genetic association and did not measure gene-gene or gene-environment interactions because not all included studies contained this information. Second, the number of studies included was relatively small, especially for studies in non-Asian populations. Therefore, the association of MIR499A rs3746444 polymorphism was associated with an elevated breast cancer susceptibility among Asians, but not among Caucasians. Additional studies is required to better clarify the clinical impact of this genetic association. Future work which takes into account gene-gene, gene-environment and gene-nutrient interactions is warranted for a more precise evidence and to further elucidate the underlying mechanism of breast cancer susceptibility.In conclusion, we provided quantitative evidence that


These results were consistent in multiple sensitivity and subgroup analyses. Significant heterogeneity was observed in the meta-analysis; however, no significant publication bias was detected. Conclusion: TCA medication may indicate an increased risk of fracture. TCA should be prescribed with caution in the clinic.Background: Research findings regarding the association between tricyclic antidepressant (TCA) treatment and the risk of fracture are not consistent; we aimed to assess whether people who take TCAs are at an increased fracture risk. Methods: Relevant studies published through June 2020 were identified through database searches of MEDLINE, EMBASE, Scopus, PsycINFO, ISI Web of Science, WorldCat Dissertations and Theses from each database’s inception, as well as through manual searches of relevant reference lists. Two researchers independently performed literature searches, study selection, data abstraction and study appraisal by using a standardized protocol. Frequentist and Bayesian hierarchical random-effects models were used for the analysis. The heterogeneity and publication bias were evaluated in this study. Results: Eight studies met the inclusion criteria. Overall, TCA use was associated with a significantly increased risk of fracture in both the frequentist approach (Risk Ratio (RR), 1.23; 95% CI, 1.06–1.42;  With the rapid growth of aging populations on a global scale, osteoporotic fractures have become a critical public health concern worldwide. Globally, around 1.6 million hip fractures occur each year , especiaOn the other hand, depression or depressive disorder is a prevalent mental illness, one that affects 17.3 million adults in the United States alone . A widelThis meta-analysis aimed to quantitatively assess all eligible cohort studies that examined the effect of TCA use on fracture risk and to obtain a more comprehensive assessment and a more precise and accurate estimate about this effect. Besides the classical frequentist method, we also employed a Bayesian approach for this meta-analysis research, as Bayesian meta-analysis uses the probabilistic method and can help with clinically relevant decision-making when one is confronted with uncertainty.We conducted this meta-analysis research using the guidelines of Meta-analysis of Observational Studies in Epidemiology (MOOSE) ,33. The We conducted a comprehensive literature search of MEDLINE (from 1946) using OVID. Without language restrictions, we used the search terms tricyclic antidepressants, antidepressant, amitriptyline, nortriptyline, protriptyline, imipramine, desipramine, doxepin, trimipramine, fractures, osteoporosis, osteopenia, bone density and bone  human subjects, (2) fracture or bone mineral density (BMD) as the outcome and (3) TCA use as exposure. Each title and abstract of each article retrieved from the electronic search was independently reviewed by two investigators (Q.W. and Y.X.). Only those citations that both reviewers deemed irrelevant were excluded. References with a disagreement between the two reviewers were included for a further full review.In the second phase of the study selection, the full content of each study obtained during the screening stage was reviewed and assessed. We included cohort studies that reported data on the subjects who had used TCAs and on the participants who were not exposed to antidepressants. The outcome variables were reported as either fractures, BMD or both. No clinical trial studies were found to be eligible. Cross-sectional studies were excluded from this meta-analysis. Using prespecified selection criteria and assessment protocols, two investigators (Q.W. and Y.X.) independently assessed the full content of each article in English and Chinese. Articles in other languages were reviewed and evaluated by additional investigators who had the corresponding multilingual expertise, using the same criteria and assessment protocol. Areas of disagreement or uncertainty were resolved by consensus. We only included studies that reported the hazard ratio (HR) or risk ratio (RR) of fractures, or/and the BMD change associated with TCA use. HR is broadly equivalent to RR ; thus, wWe evaluated the methodological quality of the included studies by using the Newcastle-Ottawa Scale . As recoWe used the data abstraction form modified from our previous meta-analysis . Two inviθ) and known within-study variance iφ). The set of iθ across the original studies was also assumed to follow a Gaussian distribution, with an unknown mean (μ) and across-study variance (τ2), where μ was the estimate of the overall log RR and τ2 was a measure of the between-study variation. The prior distribution of τ2 was assumed to follow an improper uniform distribution, and the prior distribution for τ2 was assumed to be non-informative. The probabilities that TCA use increases fracture risk by more than 0%, 10% or 20% were estimated and reported.We used the confounder-adjusted RR to measure the association between TCA use and fracture risk. HR was considered equivalent to RR. For studies that reported the estimates by subgroups only, the overall effect size was estimated by a meta-analysis of the reported subgroup’s estimates. To normalize the data distribution and to stabilize the variance, we transformed HRs or RRs into their natural logarithms  for the 2 were employed to assess heterogeneity [To assess the robustness of our estimates, we conducted several prespecified sensitivity analyses. The effects of TCAs on fracture risk were calculated with different inclusion criteria, including using TCA as the primary exposure, using osteoporotic fractures as the outcome, and studies focusing on people younger than 65 years old. The Cochran Q statistic  and the ogeneity . We applTo determine whether demographic and clinical variables modified the effect of TCAs on fracture risk, we conducted several prespecified subgroup analyses. These variables included the anatomical site of the fracture, adjustment of fracture-related confounders , year of publication and study location. As most studies did not specify race/ethnicity and others used multiethnic populations, a subgroup analysis for race/ethnicity was not performed. We conducted a cumulative meta-analysis by performing sequential random-effects pooling, beginning with the earliest qualified report. Each subsequent meta-analysis summarized all eligible reports in the preceding years. To demonstrate the effect of adding reports on the pooled effect size, we presented results chronologically in a forest plot. A multivariate meta-regression analysis was not performed because the number of eligible studies was small, and because some key variables, such as gender and race/ethnicity, were not available in some original reports.p-value of 0.05 or less was considered statistical significant.To examine the potential for publication bias, we constructed a funnel plot by plotting RRs against their standard errors . We alsoThe study selection flow is illustrated in p < 0.01), and the Higgins I2 index was 82%.The estimated fracture risk associated with TCA use changed little when studies had different inclusion criteria . For exap > 0.15). A moderate to high heterogeneity was observed in most of these subgroup analyses.The present meta-analysis demonstrated that TCA use is associated with a significantly increased risk of fracture in cohort studies. The association between TCA use and fracture risk was consistent and significant in both frequentist and Bayesian meta-analyses, as well as in all sensitivity analyses and multiple subgroup analyses, which suggested that our findings were robust. Although the fracture risk associated with TCA use found in this meta-analysis was moderate, the estimated absolute risk differences associated with TCA medications could be substantial, given that depression is a prevalent mental illness on a global scale, and that TCAs are still the major prescribed medicines for depressions that have not responded to safer agents or for which safer alternatives are not available . Thus thThe present study results are consistent with our previous meta-analysis, according to which TCA medication may convey an increased risk of fracture . Our priThe findings from the Bayesian meta-analysis were consistent with the results generated from the classical meta-analysis approach. Moreover, the Bayesian meta-analysis provided the probabilities that TCA use increases fracture risk by 10% and 20%. The calculated probabilities at different risk levels offer essential information to physicians in order to help them make clinical intervention decisions regarding the prescription of TCAs for patients. Heretofore, such information was impossible to generate when using a conventional meta-analysis methodology alone.We observed some variations in the estimated fracture risk associated with TCA use in the subgroup meta-analysis. Fracture risk associated with TCA use in studies that focused on hip/femur fracture outcomes was significantly higher than that in the nonvertebral fracture group. Since hip fracture is more likely to be influenced by falls ,47, TCAsThe underlying mechanism of how TCA medication influences fracture risk remains unclear. The increased risk of fracture might be caused by bone loss and a higher propensity to fall. Prior research found a significant association between antidepressant use and an increased risk of falls among the older population ,49. TCA The adverse effects described above make TCAs an inappropriate psychotropic drug for older patients. The Beers Criteria for Potentially Inappropriate Medication Use in Older Adults and Screening Tool of Older Persons’ potentially inappropriate Prescriptions (STOPP) ,54 recomn = 8), we were not able to perform a multivariate meta-regression analysis to further examine the sources of heterogeneity observed in this meta-analysis. Nonetheless, as described above, we conducted numerous subgroup analyses to determine the variations of the estimated risk by key variables and risk factors. Heterogeneity may be partially caused by subgroup differences in the anatomical site, year of publication, and the study sample and location, as well as by the adjustment for BMD, smoking or osteoporosis. Second, we were not able to assess the risks of falls in this meta-analysis because few eligible original studies accounted for the risk factor of falls. Nonetheless, falls remain an essential potential etiology in the association between TCA use and the risk of fracture. Third, we were not able to assess the effects of different doses of TCA on fracture risk, as the TCA dose was not available in most original studies of this meta-analysis. Fourth, other medications, including glucocorticoids [There are several limitations associated with our meta-analysis. First, because of the limited number of studies that met the inclusion criteria (rticoids ,57 and articoids , were alrticoids ,60,61. TOur study suggests that the use of TCAs is associated with an increased risk of fracture. The increased risk associated with TCA use is moderate but may have a substantial clinical impact. Nonetheless, further extensive cohort investigations are warranted, given the variance observed in the subgroup analysis and the unstable estimate found in the cumulative meta-analysis. TCA should be prescribed with caution in the clinic. Fracture risk should be monitored in patients with TCA treatment.


BACKGROUND: Previous studies have demonstrated some associations between dietary vitamin A intake and ovarian cancer risk with an inconsistent relationship. We therefore performed the present study to further explore the association between them.METHODS: Databases of PubMed, Embase, and Web of Science were retrieved up to September 1, 2019. Summarized relative risk (RR) with corresponding 95% confidence intervals (CI) were calculated. Stata 14.0 software was used for data analysis.RESULTS: Fifteen articles involving 4882 cases and 443,179 participants were included in this meta-analysis. A positive association between dietary vitamin A intake and ovarian cancer risk was found . Significant association was also found in case–control studies , but not in cohort studies. When we performed the analysis between ovarian cancer risk and geographic locations, we found an inverse association in North American populations , instead of other populations.CONCLUSIONS: In summary, findings from the present study suggested that higher dietary intake of vitamin A may contribute to the lower development of ovarian cancer, especially among North Americans. Ovarian cancer is the most deadly gynecological cancer. The American Cancer Society had estimated that there were 22,240 new cases developing in ovarian cancer and 14,070 ovarian cancer cases deaths in 2018 . However*’ OR ‘retinol’ combined with ‘ovarian cancer’ OR ‘ovarian tumor’. The bibliographies of the collected studies and relevant reviews were retrospectively assessed to identify additional articles. All the studies enrolled using this strategy was checked independently by two authors.Three electronic databases  were searched for relevant studies that investigated the association between dietary vitamin A intake and risk of ovarian cancer from inception up to September 1, 2019. The following search terms were used: ‘vitamin A’ OR ‘vitaminStudies were included based on the following criteria: (1) patients were diagnosed of ovarian cancer; (2) observational studies; (3) the interested association was about dietary vitamin A or retinol intake and ovarian cancer; (4) available relative risk (RR) and 95% confidence interval (CI) for ovarian cancer.The following exclusion criteria were used: (1) reviews or meetings or abstracts or letter to the editors; (2) overlap articles or populations; (3) animal studies; (4) no available data of RR and 95%CI; (5) vitamin supplement.Two researchers independently reviewed and extracted relevant information from all included studies. These pieces of information included: name of the first author, publishing date, country, ages, study types, vitamin types, sample sizes of the cases and controls, RR and 95%CI for ovarian cancer, adjustment for covariates. The disagreements with these two researchers were resolved by discussion and consensus. The Newcastle–Ottawa Quality Assessment Scale was used to assess the quality of the included studies .Q and I2 test were performed to analysis the heterogeneity of the studies that included in this meta-analysis [P value less than 0.05 was considered statistically significant.Data were summarized using a random-effects model for combined RR with its 95%CI . A Q andanalysis . Furtheranalysis . Statistn = 1); did not report RR (n = 8); reviews (n = 8); vitamin A supplement (n = 2); animal studies (n = 4); serum studies (n = 6); assessment of survival rate of ovarian cancer (n = 1)). Finally, 15 articles [After searching the pre-defined three databases, we got 3272 citations. Through removing the duplicated literatures among databases and those obviously did not meet the criteria by reading the title and abstract, 45 full texts of the papers were downloaded for the further screening. Thirty articles were further excluded due to some reasons   , when coP = 0.318) and funnel plot  was significantly associated with this high heterogeneity. When we performed the subgroup analysis by geographic locations, the I2 was reduced to 38.4% in North American populations and 0.0% in European populations. The result in subgroup of North American populations was consistent with the whole pooled result, while 12 of the 15 included studies were from North America.We found significant between-study heterogeneity in the whole pooled results of dietary vitamin A intake and ovarian cancer risk. Between-study heterogeneity in the meta-analysis is common, and it is an essential component to explore the heterogeneity existed in the between-study. Meta-regression was used to explore the causes of heterogeneity for covariates of publication year, vitamins type, study design, geographic locations and number of cases. Results from meta-regression suggested that geographic locations (However, several limitations should be attention. First, only English language articles were included, which may omit other languages studies. Furthermore, we only searched the articles which had been published in the journal, and did not be able including the meeting articles and some unpublished articles. However, we did not detect any publication bias, suggested our results were stable. Second, 10 of the 15 studies were case–control studies and only 5 were cohort studies. The selection bias, recall bias and some other confounding factors cannot be excluded; for example, some subjects may change their dietary vitamin A intake after the baseline assessment. However, case–control design was a very important epidemiological approach in the observational study. Therefore, it is requirement for evidence from prospective cohort studies. Third, 12 of the 15 studies were from North America, and the result was consistent with the whole pooled result. However, only 2 studies were from Europe and 1 study from Asia. We did not obtain an inverse association between dietary vitamin A intake and the risk of ovarian cancer in Europeans and Asians. Therefore, the result in the present study was more suitable for North America, but not for any populations else. Thus, more studies conducted in some other populations, instead of North Americans, are warranted to further explore the relationship between dietary vitamin A intake and the risk of ovarian cancer.In summary, our results concluded that dietary intake of vitamin A may contribute to the lower development of ovarian cancer, especially among North Americans. As some limitations existed in our analysis, large scale studies with detailed amount of dietary vitamin A intake are needed to verify our results.


Neonatal hyperbilirubinemia (NNH) is a common disease in newborns. This research study aimed to assess the associations between uridine diphospho-glucuronate-glucuronosyltransferase 1A1  polymorphisms and NNH risk.We searched PubMed, the Cochrane Library, and the Embase electronic databases. All published eligible studies before July 1, 2019, were searched for this meta-analysis.P = 0.039; GT + GG vs TT: OR = 1.331, 95% CI: 1.055–1.679, P = 0.016). Although not statistically significant, the data showed that c.3279 T > G had a tendency to be associated with NNH under the allele model and GG vs GT + TT in the overall population .We identified 7 independent studies including 1560 cases. The data showed that in the general population, compared with the GT + GG vs TT and GG vs TT, c.-3279 T > G (rs4124874) was significantly related to a higher NNH risk  polymorphism increased susceptibility to NNH, especially for the comparison of GT + GG vs TT and GG vs TT. In the future, we can use homozygous state of the UGT1A1 gene c.-3279 T > G (rs4124874) polymorphism for the diagnosis and screening of molecular biomarkers in NNH patients.The online version contains supplementary material available at 10.1186/s12881-020-01155-2. Neonatal hyperbilirubinemia (NNH) is complex and involves multiple environmental and genetic risk factors –4. ImmatThe uridine diphospho-glucuronate-glucuronosyltransferase 1A1 (UGT1A1) gene encodes bilirubin UDP glucuronosyltransferase, which is the enzyme responsible for bilirubin glucuronidation . RecentlIn recent years, many studies have shown the role of c.-3279 T > G gene polymorphisms in NNH , 13–16. The following keywords were used to research the articles in the Cochrane Library, PubMed and Embase electronic databases: “UGT1A1” OR “uridine diphospho-glucuronate-glucuronosyltransferase 1A1” OR “c.-3279 T > G” AND “polymorphism” OR “variant” OR “allele” OR “genotype” OR “neonatal hyperbilirubinemia” OR “neonatal jaundice” OR “hyperbilirubinemia”. All eligible studies were published before July 1, 2019. In addition, the reference lists were manually searched for additional studies. When multiple articles contained studies of the same population, only the most recent or complete study was used in the meta-analysis. The language of the publications was limited to English.Articles had to meet the following criteria to be included in the meta-analysis: (1) The relationship between the genetic polymorphism of UGT1A1 (c.-3279 T > G) and the risk of NNH was evaluated; (2) a case-control study; and (3) the control group complies with Hardy-Weinberg equilibrium. The exclusion criteria strongly adhere to the following: (1) lack of valid raw data; (2) data had serious deviations; and (3) reproduction of publications.The data were independently evaluated by two reviewers. The difference was not resolved until the consensus of each item was reached. Information such as author name, year of publication, country of origin, ethnic origin, source of the control population, genotyping method, matching factors and adjustment factors, number of cases and controls were recorded in each item.2. A heterogeneity P value below 0.05 was considered statistically significant for heterogeneity. The subgroup analysis of ethnicity was divided into Asians, Caucasians, and Africans. The random effects model was applied when the results of the heterogeneity test were P < 0.05 or I2 > 50%; otherwise, a fixed effect model was applied. Egger’s and Begg’s tests were used to detect publication bias, and P < 0.05 was considered a publication bias. A funnel graph and Egger’s test were used to display publication bias. All control groups were in Hardy-Weinberg equilibrium. STATA software was required for all meta-analyses . There were two sides for all tests.This study was conducted independently by two researchers based on the information collected through reading relevant articles. Whether to include the studies in the meta-analysis was decided in strict accordance with the inclusion and exclusion criteria and discussion. ORs (odds ratios) and 95% CIs were used to estimate the relationships between the c.-3279 T > G gene polymorphism and NNH. To assess whether the results of the data set were heterogeneous, we used the Q test and IFigure Seven independent studies were identified, which included 1560 cases. Table P = 0.039, Fig. P = 0.016, Fig. P = 0.067; GG vs TT + GT: OR = 1.583, 95% CI: 0.947–2.647, P = 0.080).For c.-3279 T > G (rs4124874), the main results are shown in Table P = 0.03; GG vs TT + GT: OR = 8.372, 95% CI: 1.832–38.260, P = 0.006, Fig. P = 0.027; GT vs TT + GG: OR = 0.235, 95% CI: 0.095–0.584, P = 0.002).The subgroup analysis in this paper is mainly based on ethnic differences, and Asians, Caucasians and Africans were grouped separately. The prevalence of the G and T alleles among different races is shown in Supplemental Table One-way sensitivity analysis was performed on the data involved in this meta-analysis. Each single study of the meta-analysis was deleted each time to reflect the overall impact on the individual data sets, and the corresponding combined results were not substantially changed.Publication bias was assessed by Begg’s test and Egger’s test Table . A graphThis study showed that the modifiable risk of neonatal hyperbilirubinemia includes multiple genetic and clinical risk factors. Genes that can exacerbate mutations have lower transcriptional activity in the first few days of life because their liver UGT1A1 is natural and immature. UGT1A1 is regulated in a developmental manner. Its activity can increase from 0.1% at 17–30 weeks of gestation to 1% at 30–40 weeks, reaching the adult level only 14 weeks after birth . It has Moreover, a meta-analysis confirmed that the TATA box polymorphism and c.211G > A (Gly71Arg) of the UGT1A1 gene polymorphism are associated with NNH –27. TherOne disturbing problem is the analysis of population groups, which is often not very reliable in the presence of evidence of disease, which indicates that the environment and race affect genetic background differently . SimultaThis meta-analysis has some limitations. First, heterogeneity may have interfered with the results of this meta-analysis. Nonetheless, we use specific research standards to strictly perform data extraction and analysis to minimize this possibility. In addition, published studies were included in this meta-analysis, and results with nonsignificant or negative results might not be published, so the possibility of bias may be increased in publication. Finally, our results have not been adjusted. If more research data can be obtained, a more accurate analysis could be performed. Other variables can be used, including race, age, family history, environmental factors and lifestyle –32.Although there were limitations in this meta-analysis, we found that the UGT1A1 gene c.-3279 T > G (rs4124874) polymorphism is associated with NNH risk. Therefore, it can be used as a screening index for NNH. In the future, we should consider using a larger sample size to confirm our results. In addition, the effects of gene-environment and gene-gene interactions must be researched in the future. An in-depth analysis of these factors provides a more complete understanding and more reliable results of the linkage between the UGT1A1 gene c.-3279 T > G (rs4124874) polymorphism and the risk of NNH.In summary, the UGT1A1 gene c.-3279 T > G (rs4124874) polymorphism increased susceptibility to NNH, especially for the comparison of GT + GG vs TT and GG vs TT. In the future, we can use homozygous state of the UGT1A1 gene c.-3279 T > G (rs4124874) polymorphism for the diagnosis and screening of molecular biomarkers in NNH patients.Additional file 1 : Supplemental Table S1. The prevalence of G and T allele among different races


MDM2 T309G single nucleotide polymorphism (SNP) and esophageal cancer susceptibility through pooling the open published data.The aim of this study was to investigate the correlation between MDM2 T309G single nucleotide polymorphism and esophageal cancer risk were screened. Genetic phenotype data of T309G single nucleotide was extracted from the original included studies. The correlation between MDM2 T309G single nucleotide polymorphism and esophageal cancer susceptibility was demonstrated by the odds ratio (OR) and its corresponding 95% confidence interval (95% CI). Publication bias was investigated by Egger's line regression test and begg's funnel plot.By systematic searching the databases of Medline, EMBASE, CBM and CNKI, the case‐control or cohort studies related to P = 0.043), recessive  and homozygous  genetic model through random or fixed data pooling method. Both begg's and Egger's line regression test indicated no significant publication bias.After systematic searching of the relevant database, nine publications were finally included in the present study. The combined data demonstrated that the subjects with the G genotype had an increased risk of developing esophageal cancer in dominant (OR = 1.13, 95% CI: 1.00–1.27, MDM2 T309G single nucleotide polymorphism and esophageal cancer susceptibility. Individuals with G genotype may have an increased risk of developing esophageal cancer.Based on the present data, there was a significant correlation between  Esophageal cancer is one of the most diagnosed malignant carcinoma of the digestive system. According to the latest statistical data of GLOBOCAN in 2018,MDM2) locating in chromosome 12q15 encodes a nuclear‐localized E3 ubiquitin ligase. The encoded protein can promote tumor formation by targeting tumor suppressor proteins, such as p53, for proteasomal degradation. 309 T > G of MDM2 gene is a common SNP site for human beings and considered to correlate with cancer susceptibility. Chen and colleaguesMDM2 T309G single nucleotide polymorphism and esophageal cancer susceptibility. Here, we provide an updated meta‐analysis relevant to MDM2 T309G single nucleotide polymorphism and esophageal cancer susceptibility by adding new publications.Murine double minute 2 , respectively. The publication screening procedure was performed according to Cochrane's handbook. The screening results were also cross‐checked by the two aforementioned reviewers. The references of the studies included were also carefully screened in order to identify potentially suitable publications.A systematic search of the electronic databases of Medline, EMBASE, CBM and CNKI was performed using the following subject terms: MDM2 T309G single nucleotide polymorphism and esophageal cancer susceptibility; (ii) papers were published in English or Chinese; (iii) the cases were patients diagnosed with esophageal cancer by pathology or cytology; and (iv) genotyping was correct. Publication exclusion criteria was as follows: (i) Case report or literature review publications relevant to MDM2 T309G single nucleotide polymorphism and esophageal cancer susceptibility; (ii) studies published in other languages; (iii) duplicated publishing data; (iv) esophageal cancer not confirmed by pathology or cytology; and (v) the genotype of GG, TG and TT could not be directly extracted or calculated from the original studies.The publication inclusion criteria were as follows: (i) Case‐control or cohort studies relevant to MDM2 T309G distribution were also extracted and cross checked.The general information and genotyping data of each individual study was individually extracted by two reviewers (G. Shen and B. Zhu). The main information such as first author, journal, control type and Hardy‐Weinberg equilibrium of the control group were extracted from the original included studies. The genotype of MDM2 T309G single nucleotide polymorphism and esophageal cancer susceptibility was expressed by the odds ratio (OR) and 95% CI. The statistical heterogeneity was assessed by I2 test. The OR was combined through the random or fixed effect method. The publication bias was evaluated by begg's funnel plot and Egger's line regression test.Stata11.0SE was applied for data analysis. The correlation between MDM2 T309G single nucleotide polymorphism and esophageal cancer susceptibility fulfilled the inclusion criteria and were included in the meta‐analysis , the statistical heterogeneity was not statistically different with the I2 = 36.6%, P = 0.125; However, for the recessive  and homozygous genetic models , the statistical heterogeneity was statistically significant. Therefore, data was pooled through the fixed effect method in the dominant genetic model and the random effect method in the recessive and homozygous genetic models, respectively.Statistical heterogeneity of each genetic model was assessed using the IMDM2 T309G single nucleotide polymorphism and esophageal cancer susceptibility was pooled by the fixed effect model. The combined OR = 1.13 , which indicated subjects with TG or GG genotype had increased risk of developing esophageal cancer in the dominant genetic model , which demonstrated the subjects with GG genotype were more susceptible to esophageal cancer compared with the TT or TG genotype in the recessive genetic model , which indicated subjects with GG genotype had an increased risk of developing esophageal cancer in the homozygous genetic model , the OR was combined by the random effect method because of statistical heterogeneity across the nine original publications. The pooled OR = 1.34 , recessive  and homozygous  genetic models through the random or fixed method. Chen and colleagues discussed the 309 T > G SNP and esophageal cancer risk by meta‐analysis in year 2015 and found that MDM2 T309G SNP was correlated with esophageal cancer susceptibility. Compared with the previously published meta‐analysis performed by Chen et al. our study added three new publications with increased statistical power and achieved the same conclusion.In the present updated meta‐analysis, nine case‐control studies relevant to MDM2 is itself transcriptionally‐regulated by p53.MDM2 309 T > G SNP were also correlated with an increased risk of solid tumors. Luan et al. reported that MDM2 T309 G polymorphism may contribute to NSCLC susceptibility, especially in the Asian population and women.et al. found that the GG genotype of MDM2 SNP309 was significantly associated with an increased endometrial cancer risk by the meta‐analysis.MDM2 T309 G SNP affects cancer susceptibility has not yet been fully elucidated. Knappskog and colleagues found that MDM2 T309 G SNP affected cancer risk through modulation of Sp1 transcription factor binding.MDM2 may have a large impact on the activity of p53‐dependent tumor suppression.MDM2 and cancer susceptibility are not fully understood, a significant correlation between MDM2 T309 G SNP and esophageal cancer has been confirmed by our present meta‐analysis.Although the exact pathogenesis of how SNP changes However, the present study has several limitations that need to be considered. First, only studies published in English or Chinese were searched and included in the meta‐analysis. This may have limited the number of potential articles retrieved. Second, only three new studies have been added to the work compared to the previous meta‐analysis. Third, due to the significant heterogeneity across the included studies, the statistical power was limited.The authors confirm that there are no conflicts of interest.


Nonlinear association was observed between alcohol consumption and T2D risk in men (p = 0.003). Dose-wise, consuming ≤57 g/day of alcohol was not associated with the risk of T2D in this study; however, alcohol intake >57 g/day was associated with increased risk of T2D in men. Overall, the association between alcohol consumption and T2D among Asian men was J-shaped. Lifestyle recommendations for prevention of T2D should include advice on limiting alcohol intake. This trial is registered with Prospero registration: CRD 42019121073.The objective of this review was to provide a summary of the literature on the dose-response relationship between alcohol consumption and risk of type 2 diabetes (T2D) in Asian populations, particularly men. The present study was recorded in PROSPERO as CRD 42019121073. We searched the PubMed-Medline, Web of Science, and Cochrane Library for studies published in any language since the database inception to January 2019. Prospective cohort studies were included in the meta-analysis. Relative risks (RRs) and 95% confidence intervals (CIs) were calculated for random-effects models and dose-response meta-analyses. In total, 8 prospective cohort studies were included. High alcohol intake was significantly associated with increased risk of T2D (RR = 1.16, 95% CI: 1.04–1.29; Q statistic  Type 2 diabetes (T2D) and its complications pose a major threat to global health and present challenges to patients, health-care systems, and national economies . AccordiReducing alcohol consumption might also mitigate the risk of developing T2D. Previously, a meta-analysis reported that moderate consumption of alcohol might reduce the risk of T2D, while high alcohol intake might increase the risk of T2D, following a U- or J-shaped relationship for new cases of T2D –9. In adThis meta-analysis was performed according to the 2009 Preferred Reporting Items for Systematic Reviews and Meta-Analysis statement (PRISMA Checklist) . The prePubMed, Web of Science, and the Cochrane Library databases were searched for studies that reported on the relationship between alcohol consumption and T2D, published in any language, since the start of the database coverage until January 2019. The search terms included alcohol-related terms , plus diabetes-related terms , and plus a term indicative of a prospective study design . We searched previously published relevant systematic reviews and meta-analysis to identify additional studies that might have been missed in database searches –11, 17. The inclusion criteria for studies were the following: (I) participants aged >18 years; (II) a prospective design; (III) exposure was alcohol consumption; (IV) outcome was incidence of type 2 diabetes; and (V) study provided the hazard ratio (HR) or relative risks (RRs) ratio with corresponding 95% confidence intervals (CIs) or data necessary to calculate them. We excluded reviews, editorials, studies involving non-Asian populations, and letters and editorials without sufficient data. Studies of other exposures and diseases were also excluded. Studies were excluded if reported consumption could not be converted into grams per day, and any abstention category included current drinkers. As the association between alcohol consumption and T2D risk has previously been reported as nonlinear, to obtain accurate estimates in this meta-analysis, studies could only be included if consumption was reported for multiple levels, including nondrinkers. As such, studies that did not report sufficient data were excluded. Finally, case-control studies were excluded due to the risk of bias associated with unreported confounding.Two reviewers independently identified eligible studies, extracted the relevant data, and performed the quality assessment of included studies. Authors were contacted for clarification in cases where essential information was not reported in full in published articles. Data inconsistencies were resolved by discussion, in consultation with an investigator not involved in the data extraction and conversion process. The final dataset was resolved by discussion, in consultation with an investigator not involved in the data extraction and conversion process. The final dataset was established by discussion and consensus. We performed a methodological quality assessment of the included studies using the Newcastle-Ottawa Scale (NOS) , which iWe examined the relationship between alcohol intake and risk of T2D based on the effect estimates (risk ratios (RR) or hazard ratios (HR)) and the corresponding 95% confidence intervals (CI), determined for all studies. We extracted the maximally adjusted RRs or HRs and CIs. First, the overall relationship between alcohol consumption and T2D risk was estimated with a random-effects model, based on summary RRs and 95% CIs for different alcohol intake levels compared with none alcohol intake . Next, wp values of <0.10 indicative of significant heterogeneity [Random-effect models were used when there was evidence of heterogeneity. Heterogeneity between studies was investigated using the Q statistic , and we ogeneity , 26. Subogeneity . Therefoogeneity .Details of study inclusion are presented in p = 0.931; Q statistic, p < 0.001; I2 = 76%)   .pnonlinearity = 0.003)  . AlcoholSensitivity analysis showed that removing one study with high heterogeneity from the present study's analysis sample did not affect the overall effect estimates  among Asian men. Our results suggest that the deleterious effects of high intake on T2D risk are statistically significant in Asian men. Moreover, our findings suggest that low to moderate alcohol intake does not reduce the risk of T2D, whereas heavy alcohol intake is associated with increased risk of T2D; the relationship between alcohol consumption and the risk of T2D among Asian men followed a J-shape in this study.In the last decades, several meta-analysis synthesizing findings on the relationship between alcohol consumption and T2D risk have become available; however, the results remain controversial. Some of these meta-analyses have shown a J-shaped relationship between outcome and exposure in population . HoweverWith respect to the dose-response between alcohol consumption and the risk of type 2 diabetes, we found that moderate alcohol exposure had no effect on the risk of T2D among Asian men. In contrast to previous meta-analyses, moderate alcohol consumption lowered the risk of diabetes , 10, 17.The overall RRs of the studies included in this analysis were heterogeneous. The sources of heterogeneity were identified by sensitivity analysis and subgroup analysis. In the sensitivity analysis, we found that no single study significantly skewed the estimates. However, subgroup analysis suggested that higher alcohol intake was associated with age below 50 years old, residence in Japan and Korea, as well as less sample and older studies (before 2010). In addition, the NOS score ≥8 was a source of heterogeneity. The possible reasons for these associations are as follows. First, age is an important risk factor for T2D . Second,There are several strengths to our meta-analysis. First, we evaluated Asian populations, accounting for regional differences, which has not been done before. Second, we retrieved and pooled results of studies with a prospective design, eliminating selection and recall biases. Third, we performed a dose-response analysis, including a wide range of alcohol doses. Therefore, we evaluated the relationship between drinking and the risk of T2D among men.Despite these strengths, this study has several limitations. First, the models reported in the evaluated studies were adjusted for a range of variables, which may differentially affect the risk of T2D. Second, the included studies stratified alcohol consumption according to different cutoff values. Third, the self-reported level of alcohol intake during the follow-up period was likely an underestimate. Fourth, we used aggregated data instead of individual-level data, which limits us to more detailed correlation analysis and more comprehensive results.In conclusion, the results of this study suggest that higher alcohol intake significantly increases the risk of T2D. Furthermore, the dose-response analysis suggests that the association between alcohol consumption and T2D risk follows a J-shaped curve among Asian men, with no reduced T2D risk associated with alcohol intake of 0–57 g/day, and increased risk of T2D associated with alcohol intake >57 g/day. However, further study is needed based on the limitations of the current analysis.


Objective: Research has shown a possible relationship between the E670G polymorphism of the proprotein convertase subtilisin/kexin type 9 (PCSK9) gene and an increased risk of coronary artery disease (CAD). However, there is no clear consensus on the subject because of conflicting results in the literature. The current meta-analysis was performed to better elucidate the potential relationship between the PCSK9 gene E670G polymorphism and CAD.Methods: There were 5,484 subjects from 13 individual studies who were included in the current meta-analysis. The fixed- or random-effects models were used to evaluate the pooled odds ratios (ORs) and their corresponding 95% confidence intervals (CIs).Results: The current meta-analysis found a significant association between PCSK9 gene E670G polymorphism and CAD under allelic , dominant , heterozygous , and additive genetic models .Conclusions:PCSK9 gene E670G polymorphism was associated with an elevated risk of CAD, especially in the Chinese population. More specifically, carriers of the G allele carriers of the PCSK9 gene may be predisposed to developing CAD. Coronary artery disease (CAD) is an important disease threatening human health. CAD-associated morbidity and mortality have been increasing annually in China due in part to an aging population, and epidemiological studies show that dyslipidemia is an important risk factor for CAD . DyslipiProprotein convertase subtilisin/kexin type 9 (PCSK9) gene belongs to proprotein convertase family encoding the protein called neural apoptosis-regulated convertase-1 (NARC-l), a unique proprotein in the subtilisin family that plays an important role in the cholesterol metabolism. PCSK9 overexpression increases plasma LDL-C levels by downregulating LDL receptor (LDLR) expression after transcription. The PCSK9 gene E670G polymorphism is associated with the coronary atherosclerosis severity and could serve as an independent predictor of the high LDL-C levels in the future.PCSK9 gene is located in lp32.3, spans 29 kb, and contains 12 exons. It encodes NARC-l with 692 amino acid residues. NARC-l is mainly expressed in liver and intestine and incises several common sequence variations of PCSK9. These variations have great effect on the plasma cholesterol level  could enhance the hepatocytes' ability to degrade LDLR, leading to decreased elimination of plasma LDL and ultimately resulting in hypercholesterolemia and an increased risk of CAD.The E670G polymorphism involves mutation of the adenosine (A) in the 23,968th position to a guanine (G). This results in the transformation of the glutamate (E) that is normally present in the 670th position of the 12th exon to a glycine (G). The PCSK9 gene E670G polymorphism and CAD has been conducted, a clear consensus has yet to be reached. In 2016, He et al. found that PCSK9 gene E670G polymorphism was associated with the CAD risk in Hainan and three provinces located in northeast China (TPNC) population. They also found that carriers of the G allele of PCSK9 gene E670G polymorphism were more susceptible to developing CAD in both of the aforementioned regions  diagnose CAD as the coronary artery stenosis was no <50% in no less than one coronary artery measured either by coronary angiography or dual-source coronary computed tomography, (c) have a control group with a genotype consistent with Hardy–Weinberg equilibrium (HWE), and (d) were either cohort or case-control studies published officially.The following inclusion criteria had to be met by the selected studies for our meta-analysis. Studies selected must (a) assess the association of CAD and Data from the articles that fit our inclusion criteria were extracted by three authors using a standardized protocol. Two were responsible for eliminating repetitive studies, whereas the third author served as an arbitrator to resolve any divergence between them. Studies that did not meet the inclusion criteria, were published in duplicate, or supplied insufficient data were eliminated. Similar data in multiple publications by similar author group were used for only one time in the present meta-analysis. Such items as the first author's name, publication year, region, genotyping method, matching criteria, the genotype number in the CAD and control groups, and sample size of CAD and controls are listed in PCSK9 gene E670G polymorphism and CAD. The heterogeneity among the studies was calculated by using the χ2-based Q tests with statistical significance set at P < 0.05 , dominant (AG+GG vs. AA), heterozygous (AG vs. AA), and additive (G vs. A) were used. The odds ratios (ORs) and their corresponding 95% confidence intervals (CIs) were used to compare the relationship of P < 0.05 . The fixP < 0.05 . If heteP < 0.05 . The poot-test. The values of TG are analyzed as non-normal distribution. The TG parametric analysis is performed after the logarithmic conversion.The measurements of serum lipid levels including total cholesterol (TC), TG, LDL-C, and HDL-C data were expressed as mean ± standard deviation, and the comparison between groups was performed by P < 0.05 level. The potential publication bias was assessed by using the funnel plot. The funnel plot symmetry was evaluated by using Egger's linear regression test on the OR with significance set at P < 0.05 level under the allelic genetic model . –16. 4–16m or CAD .PCSK9 gene E670G polymorphism and CAD under allelic , dominant , heterozygous , and additive genetic models   –4.PCSK9 gene E670G polymorphism and CAD were more significant in the Chinese subgroup than in the non-Chinese subgroup. In the Chinese subgroup, a significant relationship under allelic , dominant , heterozygous , and additive genetic models . In the non-Chinese subgroup, there was a significant association between them under allelic  and additive genetic models , but no significant association was found under dominant  or heterozygous genetic models .When we analyzed subgroups by ethnicity, we found the relationships between the P < 0.05). In our subgroup analysis, however, no significant heterogeneity was detected under these genetic models in the non-Chinese population (P > 0.05). It suggests that ethnicity might be the heterogeneity source.In the whole population, significant heterogeneity was detected under allelic, dominant, heterozygous, and additive genetic models . Because the lipid parameters could not be found in Yang et al.   .The sensitivity analysis was performed under the allelic genetic model. Removal of any one individual study did not affect our main result, suggesting that the results are relatively stable .PCSK9 gene E670G polymorphism and CAD. More specifically, carriers of the G allele of the PCSK9 gene E670G polymorphism are predisposed at an increased risk of CAD. When we stratified the data by ethnicity, we found that individuals of Chinese descent were particularly sensitized to developing CAD by the genetic mutation (P < 0.05). In individuals of non-Chinese descent, the significant association was only found under the allelic and additive genetic models (P < 0.05). However, there are only two non-Chinese studies in the current meta-analysis. The actual effect of PCSK9 gene E670G polymorphism on CAD in the non-Chinese population is not distinct yet.Our meta-analysis found a significant association between the Pheterogeneity < 0.05), but our subgroup analysis found heterogeneity to be limited to the Chinese group (Pheterogeneity < 0.05) with no heterogeneity found in the non-Chinese subgroup (Pheterogeneity> 0.05). This suggests that ethnicity was the primary source of heterogeneity (Pheterogeneity < 0.05).Heterogeneity was detected under allelic, dominant, heterozygous, and additive genetic models in the whole population   . Becauseanalysis . In theianalysis , 22. MorPCSK9 gene E670G polymorphism and CAD. Furthermore, environmental factors such as diet, smoking, diabetes, and metabolism diseases also play significant roles in CAD development. The microeffects of many other genes relating to CAD development have yet to be fully understood   , 34. PCStibility .LDLR gene. LDLR mutations occurred in 70% of FH patients (Apo B mutations have been detected in about 2–5% of cases in northern Europe, but not in other populations generally. PCSK9 mutations that lead to the acquisition of active function only account for <5% of FH cases (PCSK9 gene may be a part of genetic background of polygenetic FH, but it is difficult to say this is the disease-causing mutation in monogenic FH with the data from the current study in PCSK9 gene E670G mutation is not pathogenic as a FH-variant according to the Dutch Lipid Clinic Network Criteria to diagnose FH in the current study (PCSK9 gene E670G mutation is an FH-causing mutation. However, other PCSK9 gene polymorphisms than E670G mutation are likely pathogenic variant. In 2003, Abifadel et al. found that PCSK9, S127R, and F216L mutations cause autosomal dominant hypercholesterolemia in French (PCSK9 gene R306S variant was pathogenic in a 14-year-old Chinese girl with FH (PCSK9 Glu32Lys, Asp129Asn, Arg215His, and Arg496Trp mutants were regarded as pathogenic or likely pathogenic variants of Japanese heterozygous FH (FH is a severe autosomal monogene dominant disease characterized by a significant increase in plasma cholesterol, xanthoxam of the skin, and early-onset coronary heart disease . The maipatients . Apo B mFH cases . In Tablnt study . It is d with FH . In 2019ygous FH .There is likely difference between the effect on LDL-C between Chinese and non-Chinese population in the PCSK9 gene E670G polymorphism was significantly associated with CAD risk, especially in the Chinese population. Persons with the G allele of PCSK9 gene E670G polymorphism may be at an increased risk of developing CAD. More studies on the relationship between PCSK9 gene E670G polymorphism and CAD should be performed to further verify this conclusion.In brief, The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.Y-yL and HW researched data. Y-yL wrote manuscript and researched data. Y-yL, H-yG, and X-xY reviewed/edited manuscript. X-zL and Y-yL contributed to discussion, reviewed and edited manuscript. Y-yL and GG researched data, contributed discussion. All authors contributed to the article and approved the submitted version.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.


Many clinical studies evaluating the relationship between metabolic syndrome and esophageal cancer yielded uncertain results. The purpose of this study is to systematically assess the relationship between metabolic syndrome and esophageal cancer.We searched clinical studies on metabolic syndrome and esophageal cancer risk in PubMed, Embase, and the Cochrane Library. Meta-analysis was conducted by RevMan 5.3 softwares.A total of four cohort studies and two case–control studies met eligibility criteria and were included in the meta-analysis. Meta-analysis using a fixed-effect model indicated that MetS was related with a higher risk of EC . Subgroup analyses grouped by pathological types showed that MetS was related with a higher risk of EAC . Subgroup analyses grouped by metabolic conditions showed hyperglycemia ,hypertension , obesity  were related with a higher risk of EAC.Overall, our meta-analysis provides high quality evidence that metabolic syndrome was related with a higher risk of EAC. Among the individual components of the metabolic syndrome, hyperglycemia, hypertension and obesity may be the key factors. Esophageal cancer (EC) represents the seventh most common malignancy worldwide. In accordance with the global cancer statistics . It is tMetabolic syndrome (MetS) comprises a group of metabolic risk factors of cardiovascular diseases characterized by obesity, hypertension, dyslipidemia, insulin resistance , 3. WithMany clinical studies evaluating the relationship between MetS and EC have been implemented, which, however, yielded uncertain results. Drahos et al.  supporteWe carried out this meta-analysis following the guidelines of PRISMA  . This was the search strategy: (“Metabolic syndrome” OR “Metabolic Syndromes” OR “Metabolic Syndrome X” OR “Insulin Resistance Syndrome X” OR “Metabolic X Syndrome” OR “Dysmetabolic Syndrome X” OR “Reaven Syndrome X” OR “Metabolic Cardiovascular Syndrome”) AND . The reference lists of highly relevant articles and reviews were screened for potentially eligible studies manually.According to the PICOS framework , studies were included if they: (1) were full-text, English-language studies; (2) were case–control, cohort or cross-sectional studied (S); (3) reported patients diagnosed with EC(P) and showed comparisons between MetS patients(I) versus those free of MetS (C); (4) reported the incidence of EC (O); (5) reported adjusted odds ratios (ORs), relative risks (RRs), or hazard ratios (HRs) and their 95% confidence intervals (CIs) directly, or sufficient data that could indirectly calculate them. Studies were excluded if they were: (1) reviews, editorials, expert opinions, comments, case reports; (2) technical, cell, animal, or cadaver experiments; or (3) abstracts or conference proceedings.Two independent reviewers (Zhang and Wu) carried out study selection, data extraction, and study quality assessment based on the predefined criteria. Any discrepancies regarding inclusion/exclusion or risk estimates were resolved by discussions between the two reviewers or judged by a third senior reviewer (Wang). We collected the following items: surname of the first author, year of publication, the country where a study population resided, study design, and characteristics of enrolled participants incorporating sample size and age, the definition of MetS, adjusted ORs and their 95% CIs, and covariates in a multivariable model.We selected the Newcastle–Ottawa Scale (NOS), a 9-star system, to assess the methodological quality of the included studies . And we 2 statistic and Cochran’s Q-test were evaluated to determine statistical heterogeneity among studies [2 ≥ 50% and P ≤ 0.1); otherwise, a fixed-effect model was employed. Sensitivity analyses was conducted by omitting one study in turn to assess its effect on the overall results [We used RevMan 5.3 software was employed for this meta-analysis. We adopted ORs and its 95% CIs as the effect quantities, which should be converted to their logarithms and standard errors (SEs) . The I2  studies . A rando results . When th results . Egger r results .During database searching, 273 relevant studies were identified from three pivotal databases and 99 duplicates were discarded. After reading the titles and abstracts, 117 ineligible studies were excluded, and the remaining 18 full texts were reviewed. Subsequently, 12 studies were excluded due to unavailable outcomes (n = 6), inappropriate exposure (n = 4), and abstract reporting (n = 2). Finally, aligning with the predefined inclusion criteria, six studies , 24–27 iThe quality of the included studies regarding the NOS scale was shown in Table 2 = 21%). Sensitivity analyses demonstrated that none of these studies had a substantial effect on the effect-size correlation, suggesting robust and reliable evidence from our analysis .Our meta-analysis with a total of 892,614 participants showed that MetS was related with a higher risk of EC. Interestingly, there was a pathological type dependent difference. Among the individual components of the MetS, hyperglycemia, hypertension and obesity may be the key factors of MetS affecting the occurrence of EAC.The relationship between MetS and EAC risk may be reasonable biologically. Epidemiologic and experimental clinical studies supports that MetS should be an critical risk factor for common tumors development . MetS maThe sensitivity analysis indicated that the study of Ko  was a maSubgroup analyses showed that there was a pathological type dependent difference in the risk of EC with MetS patients. The reasons still remain to be elucidated, but there are several potential mechanisms. The occurrence of EC has obvious regional distribution. EAC is the most common type of tissue in Western countries, and obesity is the main cause . ObesityAmong the individual components of the MetS, hyperglycemia and hypertension may be the key factors of MetS affecting the occurrence of EC. The long-term hyperglycemia environment of the body will induce dysfunction of cellular respiration, leading to enhanced anaerobic respiration. The long-term hypoxia environment of the body cells is easy to induce the mutation of normal cells, which will promote the transformation of normal cells into malignant tumor cells.On the other hand, hyperglycemia will also promote the production of a large number of free radicals, further increase the risk of mutation of normal cells , 43. At Subgroup analyses showed that there was a nonsignificant relationship between obesity and EC risk. This was different from other studies –50. So wAs far as we know, this is the first meta-analysis to assess the relationship between MetS and EC. However, there are some limitations in our analysis.First, because of the observational nature of the included studies, our findings should be interpreted as exploratory. Second, assessment methods of MetS were different among included studies, which may be a source of heterogeneity. Third, there is a lack of uniformity in confounding factors under the multivariable models, which may bias the effect-size estimates. Fourth, although subgroup analyses can explain some sources of heterogeneity, we cannot identify other putative sources of heterogeneity due to insufficient data, such as sex. Finally, due to the limited availability of published studies, especially in the subgroup analyses, the results should be treated with caution, and more high-quality RCTs were needed.Overall, our meta-analysis provides high quality evidence that MetS was related with a higher risk of EAC. Among the individual components of the MetS, hyperglycemia, hypertension and obesity may be the key factors. Patients with MetS should be the focus of EAC screening and benefit from closer monitoring. In addition, behavioral interventions including controlling blood pressure and blood glucose could potentially serve as preventative measures for EAC in patients with MetS.


GATM) polymorphism and risk of SIM.Statin-induced myopathy (SIM) is the commonest reason for discontinuation of statin therapy. The aim of this present meta-analysis is to assess the relationship between glycine amidinotransferase gene . This association remained significant in the subgroup with fibrates or niacin excluded. However, the association of rs9806699 G>A with severe SIM was not significant. In addition, another two variations at GATM, rs1719247 C>T, and rs1346268 T>C were also associated with declined risk of SIM.Six studies with 707 cases and 2321 controls were included in this meta-analysis. GATM polymorphism including rs9806699 G>A, rs1719247 C>T, and rs1346268 T>C may be protective factors of SIM. GATM rs9806699 G>A may only exert protective effect on mild SIM cases. Our meta-analysis indicates that GATM polymorphism may represent a pharmacogenomics biomarker for predicting incidence of SIM, which contributes to risk stratification and optimizing statin adherence.The online version of this article (10.1007/s00228-020-03019-3) contains supplementary material, which is available to authorized users. SLCO1B1), rs4149056 T > C, increases the risk of SIM [GATM) and risk of SIM, has also aroused researcher’s interest.Statin-induced myopathy (SIM) is the most frequently reported adverse effect of statins and is the commonest cause for discontinuation of statin therapy . Symptomk of SIM . The assGATM is located on chromosome 15q15.3 and it encodes a mitochondrial enzyme, L-arginine, glycine amidinotransferase (AGAT), which is a rate-limiting enzyme involved in the biosynthesis of creatine [GATM rs9806699 G > A polymorphism on SIM in a case-control study with SIM cases of 72 [GATM. The A allele was associated with a decline of GATM expression, leading to decline of creatine synthesis. The reduced creatine availability might affect energy metabolism of skeletal muscle cells and thus participate in the pathogenesis of SIM [GATM were found to be in linkage disequilibrium with rs9806699 [GATM polymorphism in SIM remains a much-debated topic to date. The aim of the present study is to perform a meta-analysis to explore the relationship between GATM polymorphism and SIM, which help to identify high-risk population for SIM and provide more individualized recommendation for statin users.creatine . Creatincreatine . It playes of 72 . The locs of SIM –15. Furts9806699 , and theThis meta-analysis was reported in accordance with the Meta-analysis Of Observational Studies in Epidemiology (MOOSE) guidelines . The MOOGATM polymorphism and SIM, with the following search terms: statin, gene encoding glycine amidinotransferase, GATM, polymorphism, muscle symptom, myopathy, rhabdomyolysis, and myositis. If available, Medical Subject Headings (MeSH) terms were used together with the free text terms and synonyms. In addition, the reference lists of the eligible studies were screened for additional relevant studies.We searched the MEDLINE, EMBASE, Web of Science, and Cochrane Library databases through May 2020 for case-control studies investigating the relationship between GATM polymorphism in each group, (3) sample size more than 10 patients, (4) published in English, (5) allele frequencies available from the study population. Two reviewers (M.L and F.F) independently screened the titles and abstracts of all retrieved citations to identify studies that potentially met the inclusion criteria. Thereafter, the full texts of potentially relevant studies would be obtained and independently scrutinized by two reviewers (M.L and F.F) based on the above-mentioned inclusion criteria to be finally included. Disagreements were resolved by discussing until a consensus was reached.We included studies that met the following criteria: (1) case-control studies investigating statin-induced myopathy, (2) genotyping and reporting GATM polymorphism data  at different sites), validity of the genotyping method, and Hardy–Weinberg equilibrium (HWE) condition of each group. Disagreements were resolved by discussing until a consensus was reached.Relevant data of included studies was extracted from eligible studies using a standardized form  by two reviewers (M.L and Y.Z) independently. The following information was extracted from each study: the first author’s name, country, sample size, publication year, type of statins used, presence of interacting medications, diagnosis criteria for SIM, The occurrence of SIM was studied as outcome. SIM was defined as muscle symptoms or creatine kinase levels > 3 × upper limits of normal (ULN). The diagnosis of severe SIM is according to the specific diagnosis of each included study if it is available.The Newcastle–Ottawa Scale was used to assess the quality of studies and the risk of bias . StudiesGATM polymorphism and SIM risk was reported as odds ratio (OR) with 95% confidence interval (95% CI), based on allele contrast model. Data was pooled using either random effects model or fixed effects model according to the statistical heterogeneity between studies, which was measured by Cochran Q test and the I2 statistic. Heterogeneity between studies was defined a priori as significant when I2 statistic ≥ 50% or P value of Q test < 0.05 [P value was less than 0.05. The graphic compositions were performed by R version 3.3.1 (R Core Team).The connection between t < 0.05 , 19. If t < 0.05 . PlannedGATM rs9806699, and population of SEARCH trial was genotyped for GATM rs1719247 and rs1346268. All of the six included studies had allelic information at rs9806699, while allelic information at rs1719247 and rs1346268 were available in the study of Floyd et al. and two independent population of Mangravite et al. Various statin treatment protocols were applied, and three of six studies excluded fibrates or niacin comedications [The PRISMA flow diagram of the study selection was shown in Fig. ications , 12, 13.ications –13. The GATM rs9806699 G>A had a protective effect against SIM . There was no statistically heterogeneity according to the Q test (P = 0.301) and the I2 statistic  , but not significant in the subgroup that not excluding fibrates or niacin uses .The combined data from six eligible case-control studies proved that st P = 0.01 and thGATM rs9806699 G>A was also analyzed in different populations , but not significant in the Asian subgroup . GATM rs9806699 G>A was found to be associated with decreased risk of SIM only in studies using elevation of CK levels as a necessary condition for diagnosis criteria of SIM  . Similarly, the association of rs1346268 T>C with SIM was also significant   Fig. . There wGATM rs9806699 remained significant after excluding Floyd’s study    eFig. .Publication bias, evaluated by funnel plot, was shown in Fig. GATM rs9806699 G>A was associated with decreased risk of SIM. This association remained significant in the subgroup with fibrates or niacin excluded, indicating that GATM rs9806699 G>A might be an independent protective factor for SIM. The association of rs9806699 G>A with severe SIM was not significant. Furthermore, other two SNPs of GATM, rs1719247 C>T and rs1346268 T>C were also related to reduced risk of SIM. Our meta-analysis addresses the previous controversy and clarified the effect of GATM polymorphism on SIM, which may help to better understand the underlying mechanism of SIM and contribute to individual risk stratification for statin users.Our meta-analysis showed that GATM rs9806699 G>A was observed in our present meta-analysis. This correlation was firstly reported by Mangravite et al. in 2013. By using gene expression profiling of lymphoblastoid cell lines derived from 480 participants treated with simvastatin, they identified that an eQTL for GATM, rs9806699, interacted with simvastatin exposure. GATM rs9806699 G>A was further found to be associated with decreased incidence of SIM in their study population of 72 cases and 220 controls, which were matched based on statin exposure, age, and gender [GATM rs9806699 were only significant in studies using CK levels as criteria. Cases in Floyd et al.’s [GATM variant only protect against mild but not severe SIM. Other differences in the study populations might also lead to inconsistent results.A protective effect of d gender . Howeverd gender –15. Luzud gender  only useet al.’s  study weThe association of variations at another two SNPs, rs1719247 and rs1346268, with SIM has also been investigated. In the Study of Effectiveness of Additional Reductions in Cholesterol and Homocysteine (SEARCH) with 100 myopathy cases, both of variations at rs1719247 and rs1346268 was associated with decreased risk of SIM . As a reGATM polymorphism, we performed subgroup analyses of severe SIM cases, with data from studies of Car et al. [GATM rs9806699 polymorphism and risk of severe SIM with no heterogeneity within included studies. These results indicated that GATM rs9806699 G>A might only exert protect effect against mild but not severe SIM. Regarding that all cases in the study of Mangravite et al. [GATM rs9806699 variation. Therefore, a larger study with an accurate definition of the severity of myopathy is required. Because fibrates and niacin were considered to increase incidence of SIM and they were commonly used as comedication of statin in patients with dyslipidemia [GATM rs9806699. That provided an explanation for our negative pooled results in subgroup that not excluding fibrates or niacin. So, this finding alerted researchers the necessity of excluding the effects of drugs like fibrates or niacin as much as possible in the future. There was a significant association between GATM rs9806699 G>A and SIM in western subgroup but not in Asian subgroup. The insufficient studies at present among Asian population might lead to the non-significant result, thus more evidence in Asian population is needed.To investigate potential factors modifying the effect of r et al. , Floyd er et al. , and sevr et al.  combinede et al.  were incipidemia , 22–24, GATM polymorphism. Second, due to the limited studies regarding the association of GATM polymorphism and SIM to date, subgroup analysis for potential influencing factors like statin types,or statin doses cannot be performed. Different races with genetic diversity may also be an influencing factor, unfortunately it is not feasible to draw a conclusion because there is not enough data available for subgroup analysis. Lack of original data of individual patient in our meta-analysis restricted further subgroup analyses and interpretation of differences between studies. However, our study is the largest sample size of meta-analysis to investigate the association between GATM polymorphisms and SIM, and also the first one to perform subgroup analysis according to the severity of myopathy and comedications. Larger sample studies with standardized genotyping methods at different SNP sites of GATM and accurate definition of SIM severity are warranted to further verify its protective effect on SIM. Third, all of the included studies are case-control studies because there is lack of prospective study to date. Therefore, a well-designed prospective study should be performed in the future, with other co-variants like demographic characteristics, statin dose, types, and period of treatment considered.Our meta-analysis has some limitations. First, owing to a lack of standard definition of SIM, there is heterogeneity in the diagnosis of SIM in included studies. Larger sample studies with widely accepted definition of SIM are needed to further verify the effect of GATM polymorphism is associated with the risk of SIM. Variations including rs9806699 G>A, rs1719247 C>T. and rs1346268 T>C may be protective factors of SIM. Subgroup analysis has shown that fibrates or niacin may mask its protective effect and should be excluded in future studies. The association of rs9806699 G>A with severe SIM become non-significant in subgroup analysis, indicating that it may only exert protective effect on mild SIM cases. Our study provides new insight into the genetic determinants of SIM and identification of at-risk population, which may help to reduce the incidence of SIM and optimize statin adherence.In conclusion, our meta-analysis has indicated that ESM 1(DOCX 1438 kb)ESM 2(XLSX 13 kb)