The deficiency of vitamin D receptor (VDR) or its ligand, vitamin D3, is linked to the development of renal diseases. The TaqI (rs731236) and ApaI (rs7975232) polymorphisms of VDR gene are widely studied for their association with renal disease risk. However, studies have largely been ambiguous.Meta-analysis was carried out to clarify the association of TaqI (2777 cases and 3522 controls) and ApaI (2440 cases and 3279 controls) polymorphisms with nephrolithiasis (NL), diabetic nephropathy (DN) and end stage renal disease (ESRD).The VDR TaqI C-allele under allele contrast was significantly associated with ESRD in both fixed effect and random effect models, and ApaI C-allele with ESRD only under fixed effect model. Cochrane Q-test showed no evidence of heterogeneity for TaqI polymorphism and a significant heterogeneity for Apa I polymorphism. No publication bias was observed for both the polymorphisms.The present meta-analysis identifies TaqI and ApaI polymorphisms of VDR gene as risk factors for renal diseases. The vitamin D3 undergoes two-step hydroxylation to form 25-hydroxy vitamin D3 (25-OHD3) and biologically active 1,25-dihydroxyvitamin D3 2D3) –6. The AThe rs7975232 (NG_008731.1:g.64978G > T) is an intronic variant predicted to influence splice site changes that might affect the translation of VDR. The frequency of this variant is high as evidenced by 734 and 16,751 homozygous mutants in 1000G and ExAC databases. The rs731236 (NG_008731.1:g.65058 T > C) variant is near the exon-intron boundary (GCTG/attg) and hence likely to influence splicing and thus might affect the translation of VDR. The frequency of this variant is lower than that of rs7975232 with 242 and 7505 homozygous mutants identified in 1000G and ExAC databases.Importantly, genetic studies examining the role of TakI and ApaI polymorphisms in the pathogeneses of NL, DN and ESRD remained ambiguous –6, 8–12.The literature retrieval was carried out using keywords: vitamin D receptor or VDR, renal disease, nephrolithiasis or urolithiasis, diabetic nephropathy, TaqI (rs731236) and ApaI (rs7975232) in PubMed, Medline and google scholar databases. All the free full texts were retrieved and wherever full text was not available, reprint request was sent to the corresponding author of the respective article. The criteria to include in the meta-analysis were: 1) availability of full text of the article, 2) inclusion of studies involving both cases and controls (either online or through reprint from the corresponding author), 3) availability of raw data on genotypes, and 4) restricting to studies published in only English language. The information related to each study such as first author, year of study, ethnic group or population studied, distribution of genotypes in cases and controls etc. was computed. The decision on the studies to be included in meta-analysis was taken by all the authors of this study.2), weight and 95% confidence interval of effect size were calculated. Cochrane Q test and I2 statistics were performed to test the heterogeneity in the association. The plot of 1/SE and Z-statistics was also used as an index to test heterogeneity. The publication bias was based on the rank correlation of SE and v. The fixed effect and random effect models were generated based on Mantel Haenszel and DerSimonian Lair’s methods, respectively. If no evidence of heterogeneity was found, fixed effect model was considered. If test heterogeneity was significant, random effect model was considered.The data computed in four columns wherein first two columns represent the number of variant alleles in cases and controls and last two columns represent the number of ancestral alleles in cases and controls. Log (odds ratio) or effect size and standard error (SE) are calculated based on these four column data. Based on these two parameters, variance  and I2 (0.00) statistics showed no evidence of heterogeneity in association. Egger’s test revealed no evidence of publication bias (p = 0.14). The VDR TaqI C-allele, under allele contrast fixed effect model, was associated with renal diseases calculated collectively for DN, ESRD and NL .  . Sensitivity analysis revealed that omitting either of the studies had no effect on overall outcome of disease risk.Cochrane Q-test  and I2 (48.3) statistics showed high-degree of heterogeneity in association. Egger’s test revealed no evidence of publication bias (p = 0.54). The fixed effect model showed positive association of VDR ApaI polymorphism with all the renal disease cases , whereas, random effect model showed null association   and no association with NL and DN.Of the 13 case-control studies (2440 cases and 3279 controls) retrieved on the association of ApaI polymorphism with renal disease (Table Deficiency of vitamin D or defective activation of VDR by its ligand, 1,25-dihydroxy vitamin D results in secondary hyperparathyroidism, angiotensin II-mediated renal damage and renal disease pathogenesis . On the The direct role of solar rays in the synthesis of vitamin D is well known. In human skin, solar rays facilitate the formation of vitamin D3 from 7-dehydrocholesterol, which is evident from the presence of higher mean serum vitamin D levels in summer than in winter . LikewisThis meta-analysis revealed the association of VDR TaqI and ApaI polymorphisms with ESRD risk. This is the first meta-analysis study to simultaneously evaluate the association of DN, NL and ESRD with renal disease risk. Ethnicity, sample size, gene-environmental interactions appear to be responsible for inconsistencies observed in the association studies examining VDR polymorphisms and renal diseases. The limitations of this meta-analysis include; exclusion of studies where raw data or full text were not accessible and one-to-one correlation between vitamin D3 profile and risk could not be established as no parallel studies were conducted.


The SNP rs3747333 and rs3747334 in Neuroligin 4X (NLGN4X) gene have been demonstrated to be associated with the susceptibility to Autism spectrum disorder , but the results are inconsistent. Therefore, a meta-analysis of eligible studies reporting the association between rs3747333 and rs3747334 and ASD was carried out to enhance the reliability of published results.A systematic literature search was performed using PubMed, Web of Science, Cochrane Library to search English articles concerning the relation between rs3747333, rs3747334 and ASD up to Sep. 21th, 2017. Summary odds ratios (OR) and 95% confidence interval (CI) were used to evaluate the risk of rs3747333, rs3747334 in the ASD. The heterogeneity and publication bias of the eligible studies were also evaluated.Six eligible studies involving 1284 subjects  were included in this meta-analysis. Overall, the results indicated that there was no significant risk elevation between rs3747333, rs3747334 variants and ASD . Furthermore, sensitivity analysis and publication bias analysis confirmed this result.In conclusion, our meta-analysis suggests that the rs3747333, rs3747334 in NLGN4X gene are not frequent causes of ASD.The online version of this article (10.1186/s12991-019-0227-5) contains supplementary material, which is available to authorized users. The gloNeuroligin 4X (NLGN4X) is a member of neuroligins expressed in the postsynaptic neurons and mediate transsynaptic signaling by interacting their ligand, neurexins . SeveralGiven the increasing attention directed to the inconsistencies results, to further clarify the inconclusive association between the two SNPs and ASD, we conducted a meta-analysis with available data published up to Sep. 21th, 2017.A systematic literature search was conducted to collect studies relevant for the current meta-analysis. First, we searched three online databases  for articles published in English on associations between the rs3747333, rs3747334 in NLGN4X and ASD up to Sep. 21th, 2017. Combinations of terms identifying the gene  and terms identifying the phenotype of interest  were entered as Medical Subject Heading terms and text words. After identifying a set of publications returned from each search, we collated the results and eliminated duplicates. We then reviewed the titles, abstracts, keywords and texts of all studies to exclude studies that were clearly irrelevant. Hand-searching was carried out to determine other potential eligible studies through scanning the references cited in the retrieved articles. Finally, we applied criteria to the set of studies to determine their inclusion or exclusion in the present meta-analysis. If the two reviewers (Hongli Sun and Ying Yang) disagreed, all the authors critically evaluated the studies to judge of the inclusion or exclusion of a certain study.The study was peer-reviewed to ensure high levels of methodological adequacy and to avoid the inevitable bias caused by dependence on investigators agreeing to provide data from unpublished studies, as suggested by the Cochrane Group .The study was adopted as “Original Research”.Assessment of the association between the rs3747333 or rs3747334 or both of them in NLGN4X and ASD.Written in English.All the eligible articles meet the inclusion criteria:The two authors independently extracted the following data: (1) first author, (2) year of publication, (3) location, (4) predominant ethnicity of participants, (5) study design, (6) number of participants, (7) diagnostic criteria. When the study did not report sufficient data to calculate effect size, we contacted corresponding authors via e-mail to request additional information. If the contact with the author was not successful, we could not include the study in the meta-analysis.P-value was less than 0.05. Sensitivity analysis was also carried out to evaluate the stability of the meta-analysis. Statistical heterogeneity among studies was evaluated by the Chi-square-based Q-test and the I2 index. I2 index < 50% and/or P-value > 0.10 for Q-test indicated a lack of heterogeneity among the studies. The fixed-effect model (the Mantel–Haenszel method) was utilized to calculate the pooled ORs when the heterogeneity was not significant among the studies. Otherwise, the random-effects model (the DerSimonian and Laird method) was employed. Publication bias was evaluated by the funnel plot. All statistical analyses were carried out with the STATA 12.0 .The association of the rs3747333, rs3747334 in NLGN4X and ASD was assessed by odds ratio (OR) with 95% confidence interval (CI). Subgroup (according to study design) analysis was evaluated by study type. Publication bias was considered present when After applying inclusion and exclusion criteria see “”, data fI-squared: I2 = 95.1%, P = 0.000). We further performed the subgroup analysis by study design and the result showed that there was no statistical significant association existed between the rs3747333, rs3747334 and the risk of ASD under random-effects model . There was substantial heterogeneity in the estimated effect sizes   was observed under the random-effects model. There was substantial heterogeneity in the estimated effect sizes  analysis showed that over 16,000 individuals with highlights a novel locus at 10q24.32 and a significant overlap with schizophrenia , which mIn summary, our meta-analysis indicates that there was no significant risk elevation between the rs3747333, rs3747334 variants and ASD based on the currently published data. However, the meta-analysis adds new information to the ongoing debate relating to the risk of NLGN4X for ASD. Further investigations of these findings are merited by well-designed and large-sample multi-centric studies.Additional file 1. Supplementary material.


MTHFR, DAOA, ARVCF) had never been implicated by a schizophrenia GWA study. In sum, the present study has provided a comprehensive summary of the current schizophrenia genetics knowledgebase and has made available all the collected data as a resource for the research community.Over 3000 candidate gene association studies have been performed to elucidate the genetic underpinnings of schizophrenia. However, a comprehensive evaluation of these studies’ findings has not been undertaken since the decommissioning of the schizophrenia gene (SzGene) database in 2011. As such, we systematically identified and carried out random-effects meta-analyses for all polymorphisms with four or more independent studies in schizophrenia along with a series of expanded meta-analyses incorporating published and unpublished genome-wide association (GWA) study data. Based on 550 meta-analyses, 11 SNPs in eight linkage disequilibrium (LD) independent loci showed Bonferroni-significant associations with schizophrenia. Expanded meta-analyses identified an additional 10 SNPs, for a total of 21 Bonferroni-significant SNPs in 14 LD-independent loci. Three of these loci ( Moreover, candidate gene studies continue to be conducted in schizophrenia, with an average of 100 published each year (Supplementary fig. Genome-wide association (GWA) studies have demonstrated that schizophrenia is a polygenic psychiatric disorder. The largest GWA study of schizophrenia to date has implicated more than 100 genetic loci associated with the disorder and future, even larger GWA studies are expected to increase the number of loci to >2002. SzGene includes more than 1700 genetic association studies representing 1008 genes and 8788 polymorphisms. Since 2011 SzGene has not been updated and is now designated as a legacy database. As such, the thousands of candidate gene studies conducted in schizophrenia since 2011 have not been comprehensively captured and incorporated into this valuable schizophrenia knowledgebase. To remedy this issue and to provide an overview of candidate gene studies, we systematically identified all genetic association studies published in the field of schizophrenia between 2011–2017 and merged this information with the studies contained in the SzGene database.In 2008, Allen et al. developed the SzGene database, a comprehensive catalogue of candidate gene studies in schizophreniaPolymorphism data were collected from the SzGene database, encompassing general information of the study , discription of participants , candidate gene name, genotyped SNP IDs and genotype distribution. The SzGene study also included family-based genetic studies, but the genotype distributions for these studies were not available in its database since these studies were not included in their meta-analyses. Thus, we gathered the counts of transmitted and non-transmitted alleles from these family studies. Furthermore, incorrect and incomplete information found in the SzGene database were revised after reviewing the original genetic study  was developed based on the key words extracted from the abstracts of 1711 studies in the SzGene database.(schizophrenia OR schizophrenic OR schizophrenics OR schizoaffective) AND  AND (“2010/01/01”[Date - Publication]: “2017/02/16”[Date - Publication])The querying strategy was applied to PubMed electronic database to identify candidate gene studies. To further improve the coverage of studies, publication information of SNPs in dbSNP149  double-reviewed these studies, and disagreements were resolved through discussion with an independent reviewer (CB). Studies were processed for data collection if they were English-language articles with polymorphisms detected in schizophrenia patients  the study was a case-control or family-based association study; (2) the study included genotype or allelic frequency data or sufficient summary statistics for meta-analysis and (3) the study was not an exon sequencing, whole-genome sequencing, or GWA study as raw data from these studies were not avaialble and the risk of overlap with the cases-control or family-based studies was assumed to be high.For case-control studies to be included in the meta-analysis, one or more of the following data types was required: (1) genotype distributions of candidate SNPs in cases and controls; (2) allelic distributions of SNPs in cases and controls and (3) statistics related to an association test . When more than one of these data types were available, we prioritized genotyped distributions of SNPs. In family-based studies, counts of transmission versus non-transmissions of the reference allele in parents-offspring trios were used in the meta-analyses. We attempted to contact the corresponding authors to request data when the required data were not available or incomplete in the full-access article and supplements.Minor allele frequency (MAF) information from the 1000 Genomes Project in African (AFR: pop_id = 16653 in dbSNP149), East Asian (pop_id = 16651), European (pop_id = 16652) and South Asian (pop_id = 16655) populations were downloaded from the dbSNP149 database  for statistical analysis. The meta-analysis was conducted using the “metafor” package and “meta” library in R. For all SNPs with four or more genetic studies, we computed an estimate of allelic effect  and standard error of allelic estimate (seAE), which are essential statistics for meta-analysis. Steps to obtain AE & seAE in case-control genetic studies were as followsCT is the number of transmissions of the reference allele from parents to schizophrenia case, and CN is the number of non-transmissions of the reference allele. The odds ratio and 95% confidence interval were obtained according to both the fix-effects and random-effects models. For random-effects models we obtained summary ORs and 95% CIs with the DerSimonian and Laird4, which utilizes weights that incorporate both the within-study and between-study variance. The meta-analysis pipeline was first applied to studies regardless of population and then repeated in Caucasian and Asian populations separately.For family-based stuides we adopted the approach for obtaining AE and seAE proposed by Kazeem and Farrall:I2 statistic to quantify between-study heterogeneity caused by differences in study design and ancestry. The I2 statistic was calculated as I2 value indicates significant heterogeneity among studies and the CI bounds calculated by the random-effects model are more reliable in the presence of high heterogeneity.We used Higgins’s 5. Studies were considered as outliers if the absolute value of studentized deleted residuals was greater than 1.96. To test the influence of each study, we conducted sensitivity tests applying a leave-one-out analysis approach, which excludes each study in turn. A study was considered an influential case if it led to considerable changes in the leave-one-out analysis, which can be determined by the DFFITS, DFBETAS and COVRATIO statistics. In addition to outlier and influential cases, we also performed sensitivity analyses, excluding studies that deviated from HWE at p < 0.05.To evaluate the impact of outlier and influential cases we employed an approach recommend by Viechtbauer and Cheung6 was used to quantify the asymmetry in the funnel plot. If publication bias was identified, the”trimfill” function in metafor with Viechtbauer’s recommended parameter7 was used to impute the missing studies in the funnel plot and to obtain the estimate of the effect size after adjustment for publication bias. According to the recommendations from Sterne et al.8, the test for publication bias was only applied to SNPs with 10 or more studies.Funnel plots analysis were performed to detect potential publication bias, and the regression test proposed by HarbordTo ensure completeness of our systematic literature search, we compared the studies identified by of our search to that of meta-analyses published between 1 January 2017, and 31 December 2017 using the following PubMed query:(((schizophrenia) AND ) AND (polymorphism OR polymorphisms OR SNP OR SNPs)) AND (“2017/01/01”[Date - Publication]: “2017/12/31”[Date - Publication])9. The Z-statistic and p-value were estimated using the inverse-variance method based on a fixed-effects model, but for SNPs with pronounced heterogeneity (I2 > 75%), the Z-statistic was computed using a random-effect model10.To increase statistical power of our analyses we performed two expanded meta-analyses that combined our candidate gene study data with GWA study data obtained from the Psychiatric Genomics Consortium (PGC) schizophrenia GWA study (i.e. PGC2), four published Asian GWA studies, and an unpublished Chinese GWA study (see ). Comparison of our literature search strategy with 21 schizophrenia genetic meta-analyses published after 2017, showed our search strategy was robust and more comprehensive than the reference meta-analyses, with exception of four meta-analyses that included GWA studies, non-English articles, and/or studies published after our inclusion date  in more than 3414 genes. After applying inclusion and exclusion criteria and quality control procedures (see online methods), 1188 publications investigating 550 SNPs in 168 genes were eligible for meta-analysis  in our meta-analysis and the PGC GWAS, the effect estimate (i.e. odds ratio) was in the same direction for 79% (n = 23) of them (Supplementary Table Sp < 9.09 × 10−5), 11 SNPs in eight linkage disequilibrium (LD) independent loci remained significant  were not located within risk loci identified in the largest schizophrenia GWA study produced by the PGC and three loci  had never been implicated by a schizophrenia GWA study.The median number of studies and combined sample size (case and controls) for the meta-analyses performed were five (interquartile range (IQR) = 4–8) and 5542 IQR = 3,198–8,709), respectively. Among the 550 SNPs subjected to meta-analyses, 82 SNPs were nominally associated with schizophrenia in Asians and/or Caucasians after adjusting for publication bias and sensitivity analyses  not identified in our candidate gene only meta-analyses  locus in a previous Asian GWA study11.In the first set of expanded meta-analyses, we focused on Asian studies only. Combined data from five Asian schizophrenia GWA studies . Two of these loci, represented by OPCML and an intergenic region (rs1602565) in chromosome 11, did not sit within one of the 108 schizophrenia risk loci identified by the PGC GWA study 12. Altogether, our meta-analyses (candidate gene only and expanded with GWA studies) identified 14 Bonferroni LD-independent loci region rs02565 in MTHFR, DAOA, ARVCF) of these loci had not been reported in a schizophrenia GWA study within the Asian or Caucasian populations. However, the association result of MTHFR, DAOA and ARVCF didn’t achieve the GWAS-significant threshold (p < 10E-08), more association analyses in expended population are needed before further functional study.Our study represents the first comprehensive update of the schizophrenia knowledgebase in nearly a decade. We identified 21 Bonferroni-significant SNPs in 14 LD-independent loci associated with schizophrenia susceptibility. Three  should be interpreted with caution because of their insufficient sample sizes and uncharacteristically large odds ratios. Moreover, neither of these SNPs was replicated in our second-round meta-analyses combined with GWAS results. The other seven genes/loci  were reported by previous GWAS, suggesting most of the candidate gene studies that identified these SNPs were motivated by prior GWAS findings. In this sense, candidate gene studies didn’t provide much new insight into the pathology of schizophrenia in addition to GWAS, which was consistent with the viewpoint summarized by Johnson et al. and Farrell et alPDE4B, TRAM1L1/NDST3, ARVCF) of the eight Bonferroni-significant loci identified in the candidate gene meta-analyses did not remain significant in the expanded meta-analyses, warranting cautious interpretation and further validation of these loci. Third, the exclusion of non-English articles and limited success at obtaining access to unpublished data from authors of otherwise eligible studies might have biased our results. Fourth, we did not employ analytical approaches commonly used in single SNP-based meta-analyses to estimate pooled effects under various genetic models  or among haplotypes, nor did we employ meta-regression techniques to adjust for potential confounding factors. These decisions were based on the data available to us for the majority of studies we identified and the difficulty of obtaining the necessary data from authors to adequately perform these additional analyses. Fifth, population stratification is a major bias that can affect genetic association results and is rarely accounted for in candidate gene association studies. Finally, the random-effects model is typically viewed as a conservative approach to meta-analyses; however, in the context of funnel plot asymmetry (i.e. effect estimates are related to standard errors) random-effects models can attenuate the influence of large studies and inflate the influence of small studies.The results of our meta-analyses are not without notable limitations. First, our quality control procedures attempted to identify studies that had obvious overlapping samples. However, we cannot guarantee duplicate samples were not included and as such estimated pooled effect sizes could be negatively or positively biased. Second, several loci identified in the candidate gene meta-analyses had larger than would be expected pooled odds ratios  due to the limited sample sizes available. In fact, three (2. Although candidate gene studies were an essential contributor to schizophrenia genetics a decade ago, our study findings suggest that these studies have had a modest contribution to the understanding of schizophrenia since the emergence of GWAS technology. Despite these findings, traditional candidate gene studies will likely continue but we anticipate that these studies will be used more frequently to follow-up GWAS implicated genetic variants within deeply phenotyped individuals from which greater insights on the clinical and functional relevance of these variants may be gained.In sum, we have provided the first comprehensive update of the schizophrenia genetics knowledgebase since the decommissioning of the SZGene database in 2011Supplementary methodsFigure S1Figure S2Figure S3Supplementary Data S2Supplementary Data S3Supplementary Data S4Table S1Table S2Table S3Table S4Table S5Table S6Table S7Table S8Table S9Table S10Table S11Table S12Table S13Table S14Table S15Table S16Supplementary Dataset


Abdominal aortic aneurysm (AAA) represents an important public health problem with a prevalence between 1.3% and 12.5%. Several population-based randomized trials have evaluated ultrasound screening for AAA providing evidence of a reduction in aneurysm-related mortality in the screened population. The aim of our study was to perform a systematic review and meta-analysis of the risk factors for AAA. We conducted a systematic review of observational studies and we performed a meta-analysis that evaluated the following risk factors: gender, smoking habits, hypertension, coronary artery disease and family history of AAA. Respect to a previous a meta-analysis we added the funnel plot to examine the effect sizes estimated from individual studies as measure of their precision; sensitivity analysis to check the stability of study findings and estimate how the overall effect size would be modified by removal of one study; cumulative analysis to evaluate the trend between studies in relation to publication year. Abdominal aortic aneurysm prevalence is higher in smokers and in males. On the other hand, while diabetes is a risk factor for many cardiovascular diseases, it is not a risk factor for AAA. In addition, it is important to underline that all countries, where AAA screening was set up, had high income level and the majority belong to Western Europe . Abdominal aortic aneurysm screening is fundamental for public health. It could avoid deaths, ruptures, and emergency surgical interventions if abdominal aortic aneurysm was diagnosed early in the population target for screening. Abdominal aortic aneurysm (AAA) is defined as a permanent dilation of the abdominal aorta, with a diameter of 3 cm or more , that geAbdominal aortic aneurysm represents an important public health problem with a prevalence between 1.3% and 12.5% in males, and between 0.0% and 5.2% in females . AbdominDiabetes is a risk factor for many cardiovascular diseases, it is not a risk factor for AAA. On the contrary, it is negatively associated with AAA. This evidence could be associated to therapy with metformin ,17,18,19Several population-based randomized trials have evaluated ultrasound screening for AAA providing evidence of a reduction in aneurysm-related mortality in the screened population ,23,24,25Furthermore, it has been recently shown that the implementation of a screening system reduces not only costs, but has many benefits in terms of life expectancy . Based oClinicaltrials.gov, Web of Science, and Cochrane Library databases up to 30 June 2018. The search strategy was conducted using the following terms: Abdominal Aortic Aneurysms OR Aneurysms, Abdominal Aortic OR Aortic Aneurysms, Abdominal OR Abdominal Aortic Aneurysm OR Aneurysm, Abdominal Aortic AND Screening OR Mass Screenings OR Screening, Mass OR Screenings, Mass OR Screenings AND Factor, Risk OR Factors, Risk OR Risk Factor OR Population at Risk OR Risk, Population at OR Populations at Risk OR Risk, Populations at NOT surgical repair. The period considered was 30 June 1990–1 June 2018. Only papers written in English language were considered.The papers to be included in this systematic review and meta-analysis were sought in the MEDLINE, EMBASE, Scopus, The methodology used is described in Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) Flow-Chart  30]..30].Determinants as gender, smoking habits, hypertension, diabetes mellitus, family history of AAA and CAD were considered in our Meta-Analysis. All publication years and only English language papers are included in a specific table . ObservaMales and females were considered together. The age of the population target included in our Meta-Analysis varies according to each study. Therefore, a unique range cannot be defined.Abdominal Aortic Aneurysm is a pathology diagnosed when abdominal aorta has a diameter of 3 cm or more. We included studies that evaluated the potential risk factors associated to AAA and described it above in the eligibility criteria and study design section.The papers were selected by two independent reviewers (V.F.P and L.R.); a methodologist (E.A.) resolved any disagreements.p-value was used as a measure of effect size. A random effect model was applied as a conservative approach to account for different sources of variation among studies. Heterogeneity was assessed using Q statistics and I2. Publication bias analysis was estimated using Egger’s linear regression test [Meta-Analyses were performed when there were at least four studies. Odds ratios (ORs) with 95% CI and ion test , Begg’s ion test  and Mazuion test . The triPrometa 3  was used for all statistical analysis. The total number of records identified through database searching was 1271, in addition, records identified through other sources was 13, total records 1284. A total of 15 duplicate records were removed, 1269 total records were screened, and 450 records were excluded. A total of 819 full-test records assessed for eligibility were analyzed. We excluded 779 for following reasons: 82 records because they were case reports/case series, 11 were comments, 10 were editorials, 7 clinical guidelines, 79 were systematic reviews, 13 were meta-analyses that did not include cross-sectional studies on risk factors, 577 were about other topics. A total of 40 papers were considered for systematic review  4,9,10,,60,61,62The diagnostic test used for screening was ultrasound, except for Denmark where TC scan was used .n) and AAA detection rate (%) and screening program start were reported. A total of 14 papers were considered in Meta-Analysis, 13 prevalence studies and 1 case-control hospital-based study. The selected studies for systematic review and in meta-analysis are summarized in p < 0.0001, with Q = 132.89, I2 = 90.97, p < 0.0001  , < 0.0001 . Sensiti< 0.0001 B. Cumula< 0.0001 C. Althou< 0.0001 D, the rectively) . Meta-rep = 0.018, with Q = 390.71, I2 = 98.72, p < 0.0001  , < 0.0001 . Sensiti< 0.0001 B. Cumula< 0.0001 C. Public< 0.0001 D. The abctively) . Meta-rep = 0.039, with Q = 112.34, I2 = 93.77, p < 0.0001  , < 0.0001 . Sensiti< 0.0001 B. Cumula< 0.0001 C. Public< 0.0001 D. The abctively) . Meta-rep = 0.067, with Q = 8.45, I2 = 40.85, p = 0.133 (p = 0.008) there is bias, but Begg’s and Mazumdar’s rank correlation tests (p = 0.851) do not show presence of publication bias ,  = 0.133 . Sensiti = 0.133 B. Cumula = 0.133 C. Public = 0.133 D. Accordion bias . p < 0.0001, with Q = 5.98, I2 = 33.15, p = 0.200 (p = 0.032) but, Begg’s and Mazumdar’s rank correlation tests were not statistically significant (p = 0.624) ,  = 0.200 . Sensiti = 0.200 B. Cumula = 0.200 C. Public = 0.200 D. The re= 0.624) . p = 0.01, with Q = 30.77, I2 = 90.25, p < 0.0001  , < 0.0001 . Sensiti< 0.0001 B. Cumula< 0.0001 C. Public< 0.0001 D. The abctively) . Meta-reIn this study we show the results of a systematic review and a meta-analysis of observational studies. Some Italian studies included in our systematic review have shown AAA prevalence between 1.4% and 6.2% ,19,35,36We performed a meta-analysis using the random effect according to Der Simonian and Laird for calculate the overall effect-size .Respect to the systematic review, the first important aspect that has emerged is that the organized screening is nationwide only in the UK and Sweden ,42,43,44The second important consideration is that only high-income level countries have activated AAA screening programmes as highlighted in cancer screening ,67. TherRegarding the development of AAA, smoking is the main risk factor correlated to AAA ,33,63,73Smoking is closely correlated with the diameter of the aorta; it, indeed, is bigger in smokers compared to non-smokers and also according to Al-Zahrani et al.  AAA was Current smoking is associated with occurrence of AAA at younger ages . MoreoveIn the Multicentre Aneurysm Screening Study (MASS) the benefit of quitting smoking has been shown and this benefit leads to decrease of aortic rupture .Some studies that include females show that AAA prevalence is always higher compared to males ,51,54,55p = 0.039. Regard to the role of hypertension as potential risk factor for AAA some authors [According Forsdahl et al.  male gen authors ,44,60,63 authors  suggest Respect to gender, smoking habits, hypertension and family history of AAA, the data of our meta-analysis showed presence of heterogeneity. The absence of homogeneity could be due to different sample sizes among studies included in our meta-analysis. Egger linear regression test and mostly Begg’s and Mazumdar’s rank correlation tests show absence of a publication bias. The homogeneity among studies included was supported by Cochrane’s and Higgins’s tests. For such risk factors there is no publication bias. In agreement with previous studies diabetes mellitus is not statistically significant ,15,16,19Abdominal aortic aneurysm is correlated to risk factors associated to an incorrect lifestyle, such as smoking, a wrong diet, absence of regular exercise and gender. Kent et al.  found thIn addition, our systematic review showed that all countries where AAA screening was set up, were at high income level and the majority belong to Western Europe . The purpose of this meta-analysis was to provide a contribution to future research on the role of common risk factors, such as gender, smoking habits, hypertension, CAD, family history of AAA and to address AAA screening to target population at high risk.The best method of AAA screening is ultrasonography, which is cheap, accurate, safe, rapid, noninvasive, has good reproducibility and is cost-effective. In conclusion, these findings, together with continuous lengthening of average life, foreshadow a real “vascular emergency.” Prevention is a fundamental aspect of modern medicine that should be promoted and incentivized in a healthcare system that takes care not only of the illness itself but of the person, even when one is apparently in good health.


BackgroundPsoriasis is a chronic inflammatory skin disorder which may result in an increased cancer risk due to defects of immune surveillance. The relationship between psoriasis and risk of non-melanoma skin cancer (NMSC) has not yet been fully determined. The aim of this study was to update the evidence on the association between psoriasis and risk of NMSC.MethodsWe conducted an extensive literature search of publications in Pubmed, EMBASE, and Cochrane Library without restrictions on language from inception through August 2019 using predefined keywords. Eligible observational studies were selected if they assessed the risk ratio of NMSC in patients with psoriasis. Data from included studies were extracted, and meta-analysis was performed using random-effects models.ResultsSixteen cohort studies involving 16,023,503 participants published between 1999 and 2019 met inclusion criteria and were included in this systematic review. Meta-analysis demonstrated that compared with patients without psoriasis, patients with psoriasis had 1.72 times higher risk of developing NMSC . Patients with moderate to severe psoriasis had higher risk of NMSC  than those had mild psoriasis  (P for interaction<0.001). Moreover, patients with psoriasis had significantly higher risk of squamous cell carcinoma  than that of basal cell carcinoma  (P for interaction<0.001).ConclusionsCurrent evidence suggests that patients with psoriasis may have a higher risk of NMSC than psoriasis-free patients. Periodic screening for specific cancer risk is warranted in patients with psoriasis. Psoriasis is a chronic inflammatory and immune disease that affects more than 2-3% of the world's general population.6Due to the exposure to immunosuppressive agents, methotrexate, cyclosporine, and UV therapies, there seems to be increased cancer risk in patients with psoriasis.We therefore performed a systematic review and meta-analysis of observational studies investigating the association between psoriasis status and risk of NMSC. The primary aim of this systematic review and meta-analysis was to precisely gauge the nature and magnitude of this association in patients with psoriasis. We also assessed whether the severity of psoriasis can have an impact on the development of NMSC and whether psoriasis status can affect different pathologic type of NMSC.A systematic review was performed based on the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines.Supplementary Search Strategy). We did not apply language restrictions. In addition, manual search of potentially additional citations within references of the included studies, reviews and meta-analyses was also carried out for locating additional suitable studies not found by the electronic database searches. A search for unpublished literature was not applied.We conducted an extensive literature search of publications in Pubmed, EMBASE, and Cochrane Library without restrictions on language from inception through August 2019 for identifying observational studies examining the association between psoriasis and risk of NMSC. The search free text terms were “psoriasis” AND . MeSH  terms were also searched (The study inclusion criteria included (i) being population-based or hospital-based prospective or retrospective study with cohort, case-cohort or nested case-control design; report adjusted estimates of the relative risk (RR)  with corresponding 95% confidence intervals (CIs) for the association of psoriasis and NMSC risk. When the same cohort publishedmore than one article on psoriasis and NMSC risk, the most comprehensive one with the largest sample size was selected.Two reviewers (XW and QL) independently selected studies that potentially satisfied the inclusion criteria based on their titles or/and abstracts. If necessary, we would retrieve the full text articles for a more detailed assessment and eligibility for inclusion. For each included study, two reviewers (XW and QL) independently extracted related data including first author, year of publication, study country and design, study setting, cohort sample size, percent of female, measure of psoriasis, outcome ascertainment, mean/median follow-up period, adjusted covariates, and OR, HR, RR, SIR along with their associated 95% CIs. The results of the abstracted data were cross-checked. Any disagreements were resolved by discussion or contacted a senior reviewer (ZM).The Newcastle-Ottawa scale (NOS) was used to assess study quality for observational study. This scale, ranging from 0 to 9 scores assigned for participant selection, comparability and outcome (cohort) or exposure ascertainment (case-control), and the potential for confounding.Data analysis was undertaken using Stata® version 13.0 . A random-effects meta-analysis was performed to calculate the summary RRs and the 95% CIs, using the approach described by DerSimonian and Laird et.al.Figure 1 demonstrates the flow diagram for literature selection and inclusion in this meta-analysis, which generated 7759 records. Thirty-six potentially relevant citations were then identified for full text review. Finally, a total of 16 cohort studies involving 16,023,503 participants published between 1999 and 2019 met the inclusion criteria and were included in the final analysis.Table  In summary, all studies were published between 1999 and 2019, with 11 studies involved population-based cohorts, and 5 had hospital-based cohorts. Six of the studies examined populations from North America, seven studies were from European countries and three investigated Asian populations. Eight studies used regression model adjusted for major variables such as sex, age and body mass index. In terms of method for ascertainment of psoriasis and NMSC, most of the studies used a diagnosis based on International Classification of Diseases (ICD) criteria collected in the database.Figure 2, the summarized data from the included 16 studies indicated that compared with patients without psoriasis, patients with psoriasis had 1.72 times higher risk of developing NMSC , with significant heterogeneity across studies , although we had conducted multiple hypothesis testing by nine features. Though we did not observe significant associations between psoriasis and risk of NMSC for male patients  or for basal cell carcinoma , these associations should be further examined due to the small number of studies in each of the subgroup.Supplementary Table 1); most studies used population-representative subjects, had clearly reported the ascertainment of exposure, adequate length of follow-up, had a sufficient measurement of outcomes, and only one study with abstract form (27) rated low quality for lack of information. However, subgroup analysis based on the study quality yielded similar results between high quality studies  and low to moderate studies (P=0.805). We also conducted sensitivity analysis by omitting the low quality study and recalculating the others, which did not alter the result largely.The methodological quality score of the included studies was moderate to high in 88% (14/16) of the included studies based on the NOS score .Publication bias analysis indicated a little asymmetry of the funnel plot. However, Begg's rank correlation test (P=0.398) or Egger's test (P=0.324) did not suggest the asymmetry was attributed to publication bias. Thus, we used the Duval's nonparametric trim-and-fill method to adjust the pooled RR, and the result was the same with that of the primary analysis, showing the robustness of the result . We also conducted a further cumulative meta-analysis by calculating the cumulative evidence at the time of each study, which showed that the effect estimates had been consistent over time, providing an indication of the robustness of the results when adding new evidence .Sensitivity analyses were performed by omitting one study at a time and repeating the meta-analysis, showing that the omission of any study does not have a significantly influence on the overall effect  and basal cell carcinoma .13The strengths of this updated meta-analysis were presented as follows. First, we thoroughly searched the three major databases without language or publication date limits by applying comprehensive search strategies, making the risk of missing publications less possible. Second, we made rigorously literature screening and eligibility criteria and transparently reported the findings which may minimize the possibility of publication bias. Third, to the best of our knowledge, this meta-analysis included the biggest cohort and most comprehensive analysis with more than 16,023,000 individuals regarding this topic, providing the solidest evidence for the association between psoriasis and risk of NMSC. Fourth, various stratified analyses were performed according to some influential study variables covering the study design, participant characteristics, follow-up period, study quality, exposure and outcome features. The findings were generally in line with the result of the main analysis. Another important strength of our meta-analysis that we should address was that we used a conservative method to combine the risk estimates by using the random effects model. In consequence, the main analysis yielded a considerably high inter-study variance with a prediction interval ranging from 0.81 to 3.89, which indicated that the incidence of NMSC varied considerably in future studies comparing patients with psoriasis to psoriasis-free population. Therefore, considering the variable level of bias from the included observational studies, the significant heterogeneity of the meta-analysis along with the wide prediction interval, the findings of this study from observational studies may cast a shadow upon an uncertainty of the current understanding between psoriasis and risk of NMSC, though significant risk estimate was obtained from the main analysis. From this perspective, our findings of this meta-analysis underline the limitation of observational studies on association of psoriasis with NMSC risk due to a considerable inter-study heterogeneity. We consider that with further advanced methodologies in epidemiological research and RCTs, the potential effects of psoriasis with NMSC risk could be better clarified and stratified based on specific biological or pathophysiologic levels, in which aspects that future studies are warranted.There are limitations to our study. First, the IIn summary, the up-to-date evidence from observational studies indicates some extent of association between psoriasis diagnosis and NMSC risk. Due to the large heterogeneity of previously published cohort studies, we propose that in the future, further large prospective cohorts are advocated to validate the association between psoriasis and NMSC risk. Nevertheless, periodic screening for specific cancer risk is warranted in patients with psoriasis.Supplementary Search Strategy and tables.Click here for additional data file.


Risk of developing a malignancy when born premature is unknown. We hypothesised that risk of certain cancers might be increased in youth born preterm versus term. We therefore performed a systematic review and meta-analysis to evaluate the incidence of malignancy in the context of preterm birth, according to various cancer types.The study was designed per MOOSE and PRISMA guidelines. Articles were identified through November 2015. Observational studies exploring the association between childhood malignancy and birth characteristics were included. Of the 1658 records identified, 109 full text articles were evaluated for eligibility. Random effects meta-analyses were conducted on 10/26 studies retained; 95% confidence intervals were computed and adjusted following sensitivity analysis. Publication bias was evaluated using funnel plots, Begg’s and Egger’s tests.No differences in risk of primary central nervous system tumor  and neuroblastoma  were observed in individuals born <37 versus ≥37 weeks’ gestation. Preterm birth was consistently associated with hepatoblastoma , but not leukemia, astrocytoma, ependymoma, medulloblastoma, lymphoma, nephroblastoma, rhabdomyosarcoma, retinoblastoma or thyroid cancer.Children born premature may be at increased risk for hepatoblastoma but there is no strong evidence of an increased risk of primary central nervous system tumours or neuroblastoma. There is insufficient evidence to conclude whether prematurity modulates the risk of other childhood cancers. Perinatal characteristics have been described as independent risk factors for cancer development. An estimated 10% of neonates are born preterm; more than 95% of these babies live. OxidatiExperimental data suggest that enduring impacts of early oxidative stress exposure include premature arrest of cellular proliferation, rapid tissue differentiation, oxidative DNA damage, and impaired DNA repair–8. As a We therefore conducted a systematic review and meta-analysis of observational studies to determine whether preterm birth was associated with increased incidence of childhood malignancy, according to cancer type. We hypothesised that the risk of certain cancers may be increased in children and young adults born preterm versus term.We designed the study per MOOSE and PRISMA guidelines. Articles were identified from EMBASE, Medline (including In-Process & Other Non-Indexed Citations), and all EBM Reviews  by the principal investigator (KP) from the inception date of each resource to November 1, 2015. Expandable terms were used as MESH headings and keywords: cancer, malignan*, leukemia or leukaemia, lymphoma, tumor or tumour, *blastoma, *sarcoma, *carcinoma*. Any of these terms were combined with ((prem* or preterm) and birth)) and (infant or child* or juvenile). The search was limited to children (0–21 years) and writings in English or French. Reference lists were searched for additional titles . AuthorsRegarding article selection, original observational studies conducted on humans were considered if they reported the rate of childhood cancers in the context of GA at birth. Articles were excluded if the specific GA could not be ascertained, the definition of prematurity was not <37weeks GA, the method for cancer diagnosis was not stated or was not an accepted standard at the time of diagnosis, if there was a lack of term-born controls, and if all the reported cases were included in another study from our data set. Articles pertaining to testicular tumours were excluded, as a recent meta-analysis examined their association with GA.All titles and abstracts were independently reviewed for eligibility by two reviewers  who then assessed full-text articles per selection criteria. Disagreements were resolved by consensus. Data extraction and quality assessment were performed by three independent individuals . The quality of each article was objectively appraised using the Newcastle-Ottawa Scale (NOS) for Assessing the Quality of Nonrandomized Studies in Meta-Analysis.Data was extracted by three independent authors  for study methodology, publication date, cancer type, country(ies) included, year(s) of diagnosis, age at time of diagnosis, method of diagnosis, study population, matching criteria, confounders, risk estimates by GA, and the use of adjustment models. Two authors were contacted for additional information. Although these authors responded to our queries, the details that we requested were not available. Extracted data were entered into Microsoft Excel Version 1704 . The data underwent a final review by two authors  prior to analysis to assess and correct for discrepancies.Meta-analysis was performed in Stata/IC 12.1 . Analyses were performed using odds ratios (ORs) per unadjusted risk estimates reported in the primary publications for birth at < vs ≥37 weeks GA. In the absence of unadjusted risk estimates for these dichotomized GA categories, they were calculated from frequency tables provided in the manuscript or by personal correspondence with study investigators. Unadjusted values were used to mitigate the heterogeneity induced by combining risks adjusted for different characteristics and to allow for a larger sample size (multiple articles retained did not have adjusted risk estimates). However, repeating analyses using adjusted risk estimates when available altered neither the magnitude nor the direction of the summary effects.2 statistic. Finally, publication bias was evaluated by inspection of funnel plots, Egger’s and Beggs’ tests.To account for inter-study variability and increase meta-analysis power, outcomes with five or more studies were subjected to random effects meta-analyses. Random effects meta-analyses were used due to the inherent heterogeneity present between observational studies. Outcomes with four or fewer studies were assessed from a qualitative perspective only. Due to its rare occurrence, pooled analyses stratifying cancer risks by the degree of prematurity were not performed. Heterogeneity was evaluated using the IThe database search identified 1,658 records; 193 remained after screening their titles. Of these, 84 studies were excluded after reading their abstracts. We evaluated 109 full text articles for eligibility, 26 were retained for the qualitative synthesis, of which 10 were eligible for meta-analysis . We exclOverall, six studies reported on leukemias, including one from Oksuzyan et al , which dFour manuscripts combining 1,134 patients born preterm were retained to evaluate the relationship between gestational age and ALL , 35, 36.Three studies compared AML in individuals born at <37 versus ≥37 weeks, totalling 245 cases , 31, 35 Three studies assessed risks of any leukemia following preterm birth , 21, 31 Six studies evaluated the association between premature birth and primary CNS tumors , 33, 35.2 0%) or publication bias .Our results suggest that there is no significant difference in the risk of primary CNS tumors or neuroblastoma in those born preterm versus not preterm. Data was insufficient to conclude whether there is an impact on the development of leukemias, astrocytomas, ependymomas, medulloblastomas, nephroblastoma, lymphomas, rhabdomyosarcoma, retinoblastoma or thyroid cancer when born prematurely. Preterm birth may increase the risk of hepatoblastoma. Our results echo the findings of most published studies addressing this question.A recent meta-analysis evaluating the impact of preterm birth on childhood leukemias also noted no increase in the risk of ALL or all leukemias combined, but found a modest increase in the risk of AML. HoweverOur results are also concordant with an updated case control study of adolescent CNS tumors in Scandinavia. Among cUnlike a 2010 meta-analysis, our datCancer development requires a complex mix of cellular dysregulation and aberrant immune surveillance within the host to enable significant proliferation of a malignant line. Mechanisms which have been implicated in childhood cancer include abnormal growth factor signaling pathways, ionizing radiation, oxidative stress, environmental toxin exposure, viral infection, and underlying genetic mutations or syndromes –43. AlthIt is biologically plausible that the liver remains a vulnerable organ with increased susceptibility to cancer when exposed to preterm birth. The newborn liver is a highly metabolic organ responsible for numerous endocrine and exocrine functions, while undergoing maturation, 47. ThiOur study’s findings carry strong internal and external validity due to rigorous methodology, the inclusion of high-quality studies, minimization of duplicated populations, and computation of risk effects with a precise and accurate definition of preterm birth. The studies we included spanned multiple decades and 75% recruited cases into the 2000s, reducing the population bias inherent to earlier meta-analysis. Older manuscripts included individuals born primarily before 1990, many before 1980, when the incidence of premature birth was less and fewer preterm newborns survived , potentiOur use of registry-based studies limits introduction of recall bias, but may be a source of selection bias. All studies retained in our project used population-based registries, which are well established in the United States and Scandinavia. Hence, most cases were Caucasian, and all were from developed nations. Moreover, the multifactorial nature of cancer development might translate into requiring a larger sample size to detect a change in disease risk. Thus, type II error may be present, despite our performing summary analysis when at least 200 combined cases were reported.The limited sample sizes obtained for the occurrence of cancer in individuals born preterm induces another potential source of type II error. Indeed, the occurrence of malignancy among those born prior to 37 weeks gestation was rare, with hundreds to thousands of cases of the most common cancers in exposed individuals, despite millions of subjects being included in the combined studies for these cancers. Consequently, individual studies were under-powered to present the risk of cancer by degree of prematurity, as were we. For less common cancers, the order of magnitude of exposed cases was tens to hundreds, precluding quantitative analyses of these malignancies. Although the limited sample size is a limitation in determining the risk of cancer among those born preterm vs term, it may also be interpreted as a reassuring finding. If we assume that 10% of individuals are born preterm and that studies spanning decades yielded so few cases, then the absolute risk of cancer among persons born preterm is likely very small.th percentile for GA with those born at an appropriate weight for GA as a function of completed weeks of gestation at birth.As previously mentioned, multiple studies have found increased risk of cancer among those born with low birth weight, including hepatoblastoma. Since birth weight varies as a function of GA, comparing low birth weight vs. normal birth weight infants does not discriminate between the impact of potential intra uterine growth restriction and GA on cancer development. Differentiating the effect of birthweight and GA on childhood cancer is challenging due to the limited data available. Ideally, we would compare the risk of hepatoblastoma among infants born weighing less than 10Our study suggests that prematurity has no impact on the incidence of most childhood cancers. Although the risk of hepatoblastoma may be increased, its occurrence is so rare that the National Cancer Institute’s surveillance program has been unable to calculate an incidence rate since 1999, when there were 5 to 10.5 cases per million children below five years of age, 51. ThiPreterm birth does not appear to be associated with an increased risk of most childhood cancers. Birth prior to 37 weeks GA may increase the risk of hepatoblastoma. Ongoing surveillance of aging prospective cohorts would be beneficial to provide insight into cancer risks in the aging population of individuals born preterm.S1 Table(DOCX)Click here for additional data file.S1 Fig(TIFF)Click here for additional data file.S2 Fig(TIFF)Click here for additional data file.S1 File(DOC)Click here for additional data file.S2 File(DOC)Click here for additional data file.


Studies on relationship between tubal ligation and endometrial cancer have led to contradictory findings. In several studies, however, a reduced endometrial cancer risk was suggested following tubal ligation. Therefore, a systematic review and meta-analysis was conducted to examine the relationship between tubal ligation and endometrial cancer risk.In this systematic review and meta-analysis, PubMed/Medline, Web of Science, Scopus, Embase, and Google Scholar were searched for relevant studies published up to May 30th, 2018. We compared endometrial cancer risk in women with and without tubal ligation in retrieved studies.2 = 95.4%, P = 0.001). We pooled the raw number of tables cells  of eight studies. The SOR suggested that tubal ligation was significantly associated with a lower risk of endometrial cancer . Also, given the rare nature of endometrial cancer (< 5%), different effect sizes were considered as comparable measures of risk. Therefore we pooled ten studies and SOR of these studies revealed that tubal ligation was significantly associated with a lower risk of endometrial cancer . Besides that, we pooled eight studies in which adjusted effect sizes were reported and a subsequent analysis revealed that the summary estimate of adjusted odds ratio (SAOR) was significant .Two hundred nine studies were initially retrieved from the data bases. After exclusion of duplicates and studies which did not meet inclusion criteria, ten cohort and case-control studies, including 6,773,066 cases, were entered into the quantitative meta-analysis. There was 0.90% agreement between two researchers who searched and retrieved the studies. The summary OR (SOR) was reported using a random effect model. Begg’s test suggested that there was no publication bias, but a considerable heterogeneity was observed . It is recommended that studies should be designed to reveal mechanisms of this relationship. Concerning the relationship between tubal ligation and endometrial cancer, there are contradictory results.In this study we systematically reviewed all previous published studies and then we performed a meta-analysis.This study revealed a protective effect of tubal ligation on endometrial cancer risk by approximately 42%.According to GLOBOCAN worldwide cancer incidence and mortality report, there are annually 14.1 million new cases and 8.1 million deaths due to cancers , 3. In 2Tubal ligation, known as female sterilization, is a permanent contraceptive method with more than 99% effectiveness. According to 2015 World Health Organization Report, this method was used by 19% of women worldwide . HoweverThere is some evidence regarding the role of salpingectomy in reducing the risk of epithelial ovarian cancer , 9. BilaSeveral studies have concluded that tubal ligation is associated with the following outcomes: a reduction in risk and mortality of endometrial cancer , 12, a lIn this systematic review and meta-analysis, we followed the standard guideline of “Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA)”  and the Studies published in English, within the defined timeframe, with a non-randomized design  were included in the study. Review articles, letters to the editor, and commentaries were excluded. Retrieved records from the databases were entered into the Endnote reference manager (version X7) in order to categorize, manage, remove duplicates, and record titles, abstracts, and full-texts. Finally, titles and abstracts of the articles were evaluated and after removal of unrelated articles, full texts of the remaining articles were studied. In the cases of relevant articles in which the required data were not reported, an e-mail was sent to their corresponding author and the required information was collected. In order to prevent from missing the related articles the references of retrieved articles were also manually searched. All the above-mentioned steps were carried out by two independent researchers (SM and MS1) and their disagreements were resolved in consultation with other research team members.The retrieved full text of the articles was independently evaluated by two researchers (AAH and SS) and consulted with other research team members in case of disagreement. There was 90% agreement between two independent researchers and they strongly disagreed on one study. Information about authorship, date of publication, sample size, study design, number of endometrial cancer cases in each group , place of study, and studied population were extracted from the papers. These data were entered into an Excel sheet for preparation and cleaning. Raw numbers of the tables  as well as crude and adjusted effect sizes were also extracted from the papers.To check for risk of bias in papers, Newcastle-Ottawa assessment scale adapted for case-control and cohort studies (two separate checklists) was used . The scoP-value < 0.10) and I-square statistic were used to check for heterogeneity across the studies . In cases where there was significant heterogeneity in results of primary studies, the source of heterogeneity was investigated using a meta-regression method. By this method some available and potential sources of heterogeneity such as date of publication, sample size, study quality scores, and study design were checked by “metareg” command in Stata [Cochran’s Q test  were summarized and a summary OR (SOR) was reported to examine the relationship. Moreover, a sensitivity analysis was performed when required. All analyses were done using Stata software version 14 .A flow diagram of the literature search for studies that were included into the meta-analysis is presented in Fig. The specifications of the extracted studies are reported in Table 2 = 95.4%). Two other studies only reported the standardized incidence ratio (SIR) [2 = 98.6%). Thirdly, eight studies [2 = 86.5%).Mantel–Haenszel approach was used to estimate the SOR and its corresponding 95% confidence interval (95% CI). Due to presence of heterogeneity problem, a random effects model was used to pool the OR of the studies. Firstly, we pooled the raw numbers of cells  in tables of eight studies that reported these numbers. As it was presented in Fig. io (SIR)  and hazaio (SIR) . Secondlio (SIR) , 28. The studies , 22–24 ip < 0.001, I-squared (variation in OR attributable to heterogeneity) = 95.4%, estimate of between-study variance Tau-squared = 0.1627) in eight studies (with raw numbers) that were pooled together. There was also considerable heterogeneity  in 10 studies that their ORs were pooled together. Also, there was sizeable heterogeneity among studies for which we pooled their adjusted effect sizes . As a result of these revealed heterogeneities, random effect model was used to pool the effect sizes. In addition, meta-regression method was also used to find the source of heterogeneity.There was huge heterogeneity , study design (cohort and case-control) (p = 0.616), date of publication (before and after 2008) (p = 0.613) and study quality (high and moderate) (p = 0.569) on observed heterogeneity among studies.The results of meta-regression ruled out the involvement of sample size  (2 = 48.8%).As showed in Table P = 0.46). After removal of Iversen et al. study [Publication bias was checked for the relationship between tubal ligation and endometrial cancer risk and no evidence for publication bias was found . Moreover, risk of endometrial cancer in women with tubal ligation is 0.577 times lower that that of women without tubal ligation . A considerable protective effect was also revealed in some subgroups of studies. In case of adjusted analysis, the results of studies were adjusted for various confounder variables. List of these variables for each study is shown in Table Wernli et al.  suggesteThe relationship of tubal ligation and lower risk of endometrial cancer has been concluded in previous studies , 12, 25,Although the exact mechanism of tubal ligation and reduced risk of gynecological cancer is vague, there are some theories that have aimed to explain the relationship. In particular, one theory holds that occlusion of the fallopian tubes after tubal litigation acts as a physical obstacle and can obstruct passage of exfoliative cells  from the external genitalia and vagina into the peritoneal cavity, ovary, and fallopian tubes , 16, 29.Although only 5% of endometrial cancer cases occur before age 40 , chance The results of analysis in subgroups showed that the association between tubal ligation and endometrial cancer risk was significant in cohort studies as well as in studies with high quality Compared to cohort studies; however, case-control studies usually have less sample size and therefore less power to detect the relationship.One of the main concerns in meta-analysis studies is the presence of confounding variables and their inappropriate adjustment. In this study, we extracted the adjusted effect sizes and performed a meta-analysis on them. Although different confounding variables had been adjusted in individual studies, this analysis, to some extent, reduced the existing concerns. The results of SAOR were consistent with the unadjusted results and supported the protective effect of tubal ligation. However, standardized incidence ratios, hazard ratios, odds ratios, and risk ratios were considered as comparable measures of risk in this study . As inciFollowing from previous studies, databases of PubMed/Medline, Web of Science, Scopus, and Embase were searched to find the relevant studies. Google Scholar and the references of relevant full texts were also searched to prevent from missing the relevant studies. Therefore, it seems that probability of missing the eligible studies was minimized. Nonetheless, it should be noted that only English language full-texts were retrieved and there may be some non-English articles that were missed.Begg’s test suggested that there was no publication bias, but heterogeneity among the studies was meaningful. We tried to find the source of this heterogeneity by running a meta-regression, but the results of meta-regression rejected the role of sample size, study design, data of publication, and study quality on heterogeneity among the studies. Possibly, a part of this heterogeneity is attributable to presence of different populations in the studies. For instance, Winer et al.  conducteAs it was mentioned in the methods section all kinds of tubal ligation were pooled together in our analysis. But, it should be noted that there are some differences between mono-lateral and bi-lateral salpingectomy, tubal coagulation with or without cut, and tubal ligation by stitches with or without cut. However, since the kind of tubal ligation was not specified in the studies, there was no possibility for us to conduct a sub-group analysis. Nonetheless, it is recommended to conduct a study in future to check the association of different types of female sterilization with endometrial cancer.In summary, this meta-analysis revealed a protective effect for tubal ligation against endometrial cancer risk (roughly 42% reduction in risk). To make this finding more implementable, however, additional longitudinal studies with relatively large sample size and long follow-up designs are necessary. It is also recommended that future studies should aim to reveal the mechanism of this relationship as well.


PAI‐1) rs1799889 polymorphism and venous thromboembolism (VTE) were explored by many previous studies, yet the findings of these studies were conflicting.Association between plasminogen activator inhibitor‐1 . We used Review Manager to combine the results of individual studies.PAI‐1 rs1799889 polymorphism was significantly associated with VTE in Caucasians  and East Asians . Further analyses obtained similar significant associations in these with deep vein thrombosis (DVT) and these with Factor V Leiden mutation.Forty‐eight studies involving 14 806 participants were eligible for inclusion. Combined results revealed that PAI‐1 rs1799889 polymorphism may serve as one of the predisposing factors of VTE in both Caucasians and East Asians, especially in these with DVT and these with Factor V Leiden mutation.Our findings supported that  Its age‐ and sex‐adjusted incidence rate is estimated to be around 1.2‐1.4 per 1000 person‐years, and its 30‐day mortality rate could reach up to 10%.Plasminogen activator inhibitor‐1 (PAI‐1) is a serine protease inhibitor, and it is involved in the regulation of fibrinolysis and thrombosis via inhibiting biological activities of tissue plasminogen activator (t‐PA) and urokinase plasminogen activator (u‐PA).PAI‐1 rs1799889 A/G polymorphism and VTE.PAI‐1 rs1799889 A/G polymorphism and VTE. We will also perform comprehensive analyses to investigate the effects of ethnic background, type of disease, and established risk factors of VTE  on genetic association between PAI‐1 rs1799889 A/G polymorphism and VTE.In recent years, many investigations reported findings regarding potential associations between 2This meta‐analysis was written in accordance with PRISMA guideline.2.1We searched PubMed, Web of Science, Embase, and CNKI using the following key words: “plasminogen activator inhibitor‐1,” “PAI‐1,” “plasminogen activator inhibitor 1,” “PAI1,” “serpin family E member 1,” “SERPINE1,” “polymorphism,” “variant,” “variation,” “mutation,” “SNP,” “venous thromboembolism,” “VTE,” “deep vein thrombosis,” “DVT,” “pulmonary embolism,” and “PE.” The latest literature searching update was conducted in June 2019.PAI‐1 rs1799889 A/G polymorphism and VTE in human beings; (b) providing distributions of genotypes or alleles in cases and controls; (c) Full text in English or native language of the authors (Chinese) is retrievable. Studies were deemed to be ineligible for inclusion if: (a) Not about PAI‐1 rs1799889 A/G polymorphism and VTE; (b) studies that were not carried out in humans; (c) case reports or case series; (d) reviews and comments. If we found repeated publications during literature searching, only the most comprehensive study was included for analyses.To be included in this meta‐analysis, some criteria must be met: (a) About 2.2PAI‐1 rs1799889 A/G polymorphism in cases and controls. We also calculated the probability value  of Hardy‐Weinberg equilibrium.Following information was extracted by two authors: the last name of the first author and publication year, country of the principal investigator and ethnicity of study participants, type of disease, total sample size of each study, and the distribution of The authors used Newcastle‐Ottawa scale (NOS) to assess the quality of eligible studies.Two authors extracted data and assessed quality of eligible publications. The authors wrote to the leadings authors for additional information if essential information was found to be incomplete.2.3Z test was employed to assess whether PAI‐1 rs1799889 A/G polymorphism was significantly associated with VTE, with the statistical significance P level set at .05. We used I2 statistics to assess between‐study heterogeneities. We used Random‐effect models (DerSimonian‐Laird method) to combine the results if I2 is larger than 50%. Otherwise, fixed‐effect models (Mantel‐Haenszel method) were used to combine the results. We also conducted subgroup analyses by ethnicity of participants, type of disease and whether the study subjects had established risks of VTE. We examined the stability of combined results by deleting one study each time and combining the results of the remaining studies. We used funnel plots to estimate whether our combined results may be influenced by publication biases.Review Manager was used to combine the results of eligible studies. This article does not contain any studies with human participants or animals performed by any of the authors, thus ethical approval is not required.33.1One thousand eight hundred and twenty‐nine studies were identified by our comprehensive literature searching. One hundred and thirty‐three studies were retrieved for eligibility assessment after exclusion of irrelevant and duplicate articles. Another eighty‐five articles were further excluded by us because these articles did not meet the inclusion criteria that were set forth for this meta‐analysis. Totally forth‐eight studies containing 5731 cases and 9075 controls were ultimately included in this meta‐analysis  and East Asians . Further analyses revealed similar significant associations in the DVT  subgroup, yet no any positive results regarding PE were detected in this meta‐analysis (see Table PAI‐1 rs1799889 A/G polymorphism and VTE, and we found positive results in these with Factor V Leiden mutation, whereas no any significant associations were detected in these with cancer or these who recently had a major surgery operation.3.3We examined the stability of combined results by deleting one study each time and combining the results of the remaining studies. The trends of associations remained consistent in sensitivity analyses, which indicated that the combined results were statistically stable.3.4Funnels plots were employed to estimate whether our combined results may be influenced by publication biases. Funnel plots of every comparison were symmetrical, which indicated that the combined results were unlikely to be seriously impacted by overt publication biases (see Figure 4PAI‐1 rs1799889 A/G polymorphism and VTE, yet the findings were somehow conflicting. Thus, this meta‐analysis was performed by us to more comprehensively analyze relationship between rs1799889 polymorphism and VTE. The combined results demonstrated that PAI‐1 rs1799889 A/G polymorphism was significantly associated with VTE in both Caucasians and East Asians. Further analyses obtained similar positive findings in these with DVT and these with Factor V Leiden mutation. The trends of associations remained consistent in sensitivity analyses, which indicated that the combined results were stable.The A/G variant of rs1799889 polymorphism is associated with a guanosine insertion at the −675 site of the PAI.PAI‐1 rs1799889 A/G polymorphism and PAI. Unfortunately, the majority of eligible publications only focused on genetic associations, so we could not estimate genetic‐environmental interactions in this meta‐analysis.PAI‐1 polymorphisms and VTE. Nevertheless, only rs1799889 polymorphism was analyzed by us because no any other PAI‐1 polymorphisms were studied by at least two different studies. Fourthly, it should be noted that in 2014, Wang et al also performed a meta‐analysis to investigate association between rs1799889 polymorphism and VTE.PAI‐1 rs1799889 polymorphism might serve as one of the predisposing factors of VTE. Subgroup analyses by established risk factors of VTE were also further performed by us. Nevertheless, since these analyses were only based on limited number of participants, the findings of these analyses should be considered as only exploratory, and further experimental studies should try to confirm these results. Besides, more stratified analyses should also be conducted by future meta‐analyses if there are sufficient data to support additional analyses for other established risk factors of VTE. Fifthly, although we also conducted subgroup analyses by type of disease in this meta‐analysis, it is noteworthy that studies only focused on PE were scare, so the results of subgroup analyses by type of disease should also be taken as exploratory. Future studies are still needed to confirm these findings.To better understand the combined results of this meta‐analysis, some points should be considered. Firstly, the etiology of VTE is complex, so we recommend future studies to conduct haplotype analyses and investigate gene‐gene interactions to more precisely analyze the effects of genetics on disease susceptibility.PAI‐1 rs1799889 polymorphism and VTE were based on combining unadjusted findings of eligible publications due to lack of raw data.Some limitations of this meta‐analysis should also be mentioned. Firstly, the results regarding associations between polymorphisms in 5PAI‐1 rs1799889 A/G polymorphism may serve as one of the predisposing factors of VTE in both Caucasians and East Asians, especially in these with DVT and these with Factor V Leiden mutation. Further studies with larger sample sizes still need to verify our findings. Besides, given that the pathogenesis of VTE is complex, despite our comprehensive analyses, we still recommend further studies to explore gene‐gene interactions and gene‐environmental interactions in the development of VTE.In summary, the combined results of this meta‐analysis proved that The authors declare no potential conflict of interests.Guangbin Huang and Xuejun Deng conceived of the study, participated in its design. Guangbin Huang and Pan Wang conducted the systematic literature review. Tao Li performed data analyses. Guangbin Huang and Xuejun Deng drafted the manuscript. All authors have read and approved the final manuscript.Figure S1 Funnel plots.Click here for additional data file.


Hypertension is the most common chronic disease, and most important risk factor for cardiovascular disease. This meta-analysis aimed to explore the association between hepatic lipase gene (LIPC) gene -250G/A (rs2070895) and -514C/T (rs1800588) polymorphisms and the susceptibility to hypertension.P < 0.05 or I2 ≥ 50%; otherwise, the fixed effect model was selected.Published studies were searched using the PubMed, Embase and Cochrane Library databases. Newcastle-Ottawa Scale (NOS) was used to assess the quality of the included studies. Sensitivity analysis was performed using “leave one out” method. Egger’s test was used to evaluate the publication bias. The random effect model was used to calculate the pooled effect size if P = 0.38; CT vs CC, P = 0.46; TT vs CC, P = 0.38; TT vs CC + CT, P = 0.54; TT + CT vs CC, P = 0.34). Notably, the frequencies of the AA+GA genotypes of LIPC rs2070895 polymorphism were related to an increased risk of hypertension .Four eligible studies, including 2599 participants, were enrolled in the included studies from 2007 to 2014. Quality evaluation revealed that each study had high NOS scores ranged from 5 to 7. The LIPC rs1800588 polymorphism was not found to be associated with the susceptibility to hypertension under all genetic models (T vs C, The LIPC rs2070895 polymorphism was found to be related to an increased risk of hypertension. However, LIPC rs1800588 polymorphism was not associated with the susceptibility to hypertension. The LIPC rs2070895 polymorphism was related to susceptibility with hypertension.The LIPC rs1800588 polymorphism was not strongly associated with the increased risk of hypertension.The main characteristic of hypertension –3 is incHepatic lipase (LIPC), a member of the lipase family, is located on the long arm of chromosome 15 (15 q21) and has a total length of 35 kb, comprising nine exons and eight introns –10. LIPCHypertension is an important risk factor for cardiovascular disease including coronary heart disease and cerebral infarction, among others. An increasing number of studies has focused on LIPC gene and hypertension. However, many of these studies included small sample size, and showed controversial results. Thus, this meta-analysis was performed to pool the studies of the frequency distribution of LIPC rs2070895 and rs1800588 polymorphism in hypertension patients and normal individuals to determine the correlation between LIPC gene polymorphism and hypertension, as well as to provide information for guiding future clinical studies.Relevant studies were searched using the PubMed, Embase and Cochrane Library databases. The following search keywords were used: (“Hypertension” OR “hypertensive” OR “High Blood Pressure”) AND (“LIPC” OR “hepatic lipase gene” OR “rs2070895” OR “-250G/A” OR “-514C/T” OR “C-480 T” OR “rs1800588”) AND (“polymorphic*” OR “genetic” OR “variant”). The search was performed by two reviewers. The retrieval deadline was May 1, 2018 with no language restrictions. This study followed the analysis criteria of Preferred Reporting Items for Systematic Reviews and Meta-Analyses and the participants, interventions, comparison and outcome measures principle.The included studies conformed to the following criteria: (1) the studies examined the frequency distribution of LIPC -250G/A and -514C/T polymorphisms in hypertension and non-hypertension patients; (2) the studies reported precise genotype or allele frequency data in both groups; (3) the study was a case-control analysis.Additionally, the studies were excluded if: (1) the data were incomplete, and thus could not be statistically analyzed; (2) the studies were non-treatise literatures, such as reviews, letters, or comments; (3) the study was a repetitive publication or based on data used for multiple studies. In this case, only the most recent study or that with the most complete information was included.Two reviewers independently extracted relevant data from the included publications. The following information was extracted: name of first author, published year, study regions, detection method of genetic polymorphism, number of participants in hypertension and non-hypertension groups, number of each genotype with LIPC (rs2070895 and rs1800588) polymorphisms in both groups, and demographic characters including age and gender. The quality of each included study was evaluated based on the Newcastle-Ottawa Scale (NOS) recommended by the Agency for Healthcare Research and Quality of the US   and -514C/T (rs1800588) polymorphisms in hypertension patients and normal individuals to assess the relationship of LIPC gene polymorphisms and hypertension risk. The LIPC rs1800588 polymorphism did not appear to be associated with an increased risk of hypertension. Notably, the LIPC rs2070895 polymorphism frequencies were significantly increased in the hypertension group compared to in controls under the dominant model, and OR values of LIPC rs2070895 polymorphism under all models were larger than 1. Therefore, the LIPC rs2070895 polymorphism may be a risk factor for the susceptibility to hypertension.The human HL protein encoded by the LIPC gene has phospholipase A1 and triglyceride hydrolytic enzyme activities, and plays an important role in plasma lipid transport . It is wSome studies showed a strong connection between the LIPC -250G/A rs2070895) polymorphism and lipoprotein metabolism disease, such as hyperlipidemia , 34 and 95 polymoThe T allele in LIPC -514C/T polymorphism has been shown to be related to higher plasma HDL-C levels , and theHeterogeneity test results revealed a significant difference between included studies. The potential heterogeneity sources included different regional living habits, living environments, and the economic development levels. Additionally, the effects of other confounding factors such as gender and age can lead to high heterogeneity.There were some limitations in this research. First, because of incomplete data and the small sample sizes of the included studies, covariate correction and further subgroup analysis were not performed, although these factors may have confounded the results of the meta-analysis. Second, the population was not representative, as the population in Yin’s study  did not fit HWE. Third, the results of sensitivity analysis regarding the LIPC rs2070895 polymorphism in the allele, additive, and dominant models were unstable after elimination of Lin’s study, which requires further analysis. Fourth, hypertensive disorders complicating pregnancy may have affected our results.In summary, the present study showed that the frequencies of the AA+GA genotypes of the LIPC rs2070895 polymorphism are related to an increased risk of hypertension. However, the LIPC rs1800588 polymorphism was not associated with the susceptibility to hypertension. Because the related studies and sample size were limited, the conclusions of this study should be verified by high-quality studies with larger sample size.