Contrasting results exist on the association between CTLA-4 rs231775 and acute rejection in kidney transplant recipients. We herein conducted an updated systematic review with meta-analysis and trial sequential analysis (TSA) to clarify this relationship and to establish whether the current evidence is sufficient to draw firm conclusions. In addition, noteworthiness of significant pooled odds ratios (ORs) was estimated by false positive report probability (FPRP). A comprehensive search was performed through PubMed, Web of Knowledge, Cochrane Library and Open Grey up to October 2019. Fifteen independent cohorts, including a total of 5,401 kidney transplant recipients, were identified through the systematic review. Overall, no association was detected with the allelic , dominant  or the recessive  model of CTLA-4 rs231775. In each genetic model, the cumulative Z-curve in TSA crossed the futility boundary and entered the futility area. In addition, none of the significant genetic comparisons detected in the subsequent and sensitivity analyses or in previously reported meta-analyses were found to be noteworthy by FPRP. In conclusion, this study provides strong evidence that CTLA-4 rs231775 is not a clinically-relevant genetic risk determinant of acute rejection after renal transplantation. CTLA-4 exerts its immunosuppressive function through a variety of mechanisms which include competition with the co-stimulatory CD28 molecule for binding to their shared B7 ligands (CD80/CD86) on the antigen-presenting cells (APC)2, and interference with TCR-mediated signal transduction3. Given the key role of CTLA4 on regulation of allograft rejection and tolerance5, a great attention has been focused on the relationship between CTLA-4 genetic variation and graft outcome following solid organ transplantation9.Cytotoxic T lymphocyte antigen-4  is a transmembrane homodimer glycoprotein expressed by activated effector T cells (Teffs) that negatively regulates T cell-mediated immune responses11. The single-nucleotide polymorphism (SNP) rs231775 of the CTLA-4 gene, also referred as + 49A>G, causes a threonine-to-alanine substitution at codon 17 in the peptide leader sequence, and guanine at this position is related to reduced CTLA-4 protein expression12. Over the last fifteen years, several efforts have been made to evaluate the impact of CTLA-4 rs231775 on the risk of acute rejection in kidney transplant recipients, however the results differed among studies, which reported a lack of association14 or a higher risk of acute renal rejection among GG carriers16 or inconclusive findings because of low sample size19. The reasons for these conflicting and inconclusive results may be the different ethnicity, the clinical heterogeneity, the low statistical power or a combination of these factors.The human CTLA-4 gene is located on the short arm of the second chromosome (2q33) and consists of four exons encoding respectively the leader sequence peptide, an extracellular immunoglobulin like domain containing the binding site, a transmembrane domain and a cytoplasmic tail25. However, all these studies did not take into account the risk of random errors due to sparse data and multiple meta-analytic up-dates27, which often result in false positive (type-1 error) and false negative (type-2 error) findings. Given that the above-mentioned issues can be addressed by application of trial sequential analysis (TSA) to meta-analytic results28 and the recent publication of two novel primary studies30, we herein conducted an updated meta-analysis with TSA to assess reliability of the accumulated evidence on the relationship between CTLA-4 rs231775 and acute renal transplant rejection. In addition, noteworthiness of significant pooled estimates from the present and previous meta-analyses was estimated by false positive report probability (FPRP).Up to now, six meta-analyses have been conducted to clarify the role of CTLA-4 rs231775 on the risk of acute renal graft rejection31. A computerized literature search was carried out on PubMed, Web of Knowledge, Cochrane Library and Open Grey  by using the Boolean combinations of the key terms: (cytotoxic T-lymphocyte antigen 4 OR CTLA4 OR CTLA-4 OR GWAS OR genome-wide association study) AND  AND  AND rejection. Eligible studies were required to meet the following inclusion criteria: (i) investigating the association between CTLA-4 rs231775 and acute rejection (AR) in kidney transplant recipients (KTRs) (ii) reporting sufficient data for estimating an odds ratio (OR) with 95% confidence interval (CI) for the association with CTLA-4 rs231775. Exclusion criteria were: not human studies or not related to the research topics; case reports, editorials and meeting abstracts; narrative reviews, systematic reviews with or without meta-analysis; duplication of previous publications. The potentially relevant articles were then read in their entirety to assess their appropriateness for inclusion in the systematic review. Reference lists of retrieved studies were also checked to identify other potentially eligible studies. If two or more studies shared part of the same patients’ population, the one with the larger sample size or more complete data was included. The corresponding authors were contacted by e-mail when the eligible paper had insufficient information for calculation of OR and 95% CI. Studies were excluded if the corresponding author did not answer to the e-mail or was unable to provide the requested data.This systematic review was conducted in accordance with the PRISMA Statement principleshttps://www.ohri.ca/programs/clinical-epidemiology/oxford-asp), which consists of three components: (I) selection and definition of the study groups (0–4 points); (II) comparability of the cohorts (0–2 points); and (III) ascertainment of outcomes (0–3 points). Studies with a NOS score ≥ 7 out of 9 were considered of higher quality. All studies were independently analyzed by two reviewers (S.T. and S.C.) and any discrepancies in study selection, data extraction and methodological quality evaluation were resolved through consensus.From each identified study the following data were extracted: name of first author, year of publication, study location, ethnicity, mean age, male/female ratio, donor type (i.e. living or deceased), immunosuppressive drugs, criteria for diagnosis of acute rejection, number of KTRs with and without AR, timing of AR after kidney transplantation, method of CTLA-4 rs231775 genotyping, and allele/genotype counts. Methodological study quality was assessed using the Newcastle–Ottawa scale (NOS) for cohort studies (available at: https://ihg.gsf.de/cgi-bin/hw/hwa1.pl). ORs were pooled based on the allelic (G vs. A), dominant (GG/AG vs. AA) or recessive (GG vs. AG/AA) genetic contrast of CTLA-4 rs231775 by using the random-effects (DerSimonian–Laird method) model, which takes into account both within study variance and cross-study variance32. In case of lack of heterogeneity, the random effects model coincides with the fixed-effect model33. Between-study heterogeneity was tested using the Q statistic, with a p-value < 0.10 indicating the presence of significant heterogeneity among studies. Heterogeneity was also quantified by the I2 metric, with I2 values > 50% indicating high heterogeneity34. The robustness of overall estimates was verified by conducting subgroup and sensitivity analyses. The presence of publication bias or a difference between small and large studies  in the overall analyses was evaluated graphically by drawing funnel plots and statistically by means of Egger’s test35. In case of statistical evidence of funnel plot asymmetry , the ‘trim-and-fill’ method was used to adjust the overall pooled estimate for potential publication size or small study effects36. All analyses were performed using ProMeta software  and the significance of pooled ORs was set at P < 0.05. Noteworthiness of significant pooled ORs was also estimated by false positive report probability (FPRP)37, which is calculated based on the statistical power of the test, the observed P-value, and a given prior probability for the association. FPRP values were calculated at the prior probability of 0.001 (expected for a candidate gene)38 to detect ORs of 1.50 , by using the Excel spreadsheet provided by Wacholder et al.39. A significant result (P < 0.05) with an FPRP value of less than 0.2 indicated a noteworthy association.For each study, the Hardy–Weinberg Equilibrium (HWE) was calculated using the Pearson’s goodness-of-fit chi-square test implemented in the online Finetti’s program , a statistical test power of 80% (β = 0.2) and an “a priori” relative risk difference (reduction or increase) of 15%. In addition, we set the event proportion in the reference genotype or allele group as the median value across studies included in the meta-analysis, and we adjusted the required information size for study heterogeneity by applying a D2 adjustment factor39. If the cumulative Z curve crosses the trial sequential monitoring boundaries with achievement of RIS, it means that a sufficient level of evidence has been reached and further studies are unneeded. When the Z curve does not cross any of the boundaries and the RIS has not been reached, it can be concluded that more studies are required to reach a sufficient conclusion. If the cumulative Z curve crosses the futility boundaries, the conclusion of indiscrimination between two groups is accepted under the given conditions40. TSA was performed using Trial Sequential Analysis software41 version 0.9.5.10 beta (available at www.ctu.dk/tsa).Trial sequential analysis (TSA) allows to control the risk of type I  and type II  errors of conventional meta-analysis and to calculate the required information size (RIS), that is the required number of participants in a reliable and conclusive meta-analysis46. The main characteristics of the identified studies are summarized in Table 46, three45 and two studies43. The largest study30 included a cohort of European Americans  and a cohort of African Americans (n = 482). In the majority of the identified studies, patients received renal allografts from living or cadaveric donors, while in 4 studies the transplanted kidney came exclusively from a cadaveric donor18.The flow chart illustrating the overall literature selection process is illustrated in Fig. 45; it significantly deviated from HWE in three studies42, while it was not computable in two studies46 which reported genotype data as a combined group (AG/AA vs. GG). The distribution of CTLA-4 rs231775 genotypes for each included cohort and timing of AR assessment after kidney transplantation is shown in Table 45 the association of CTLA-4 rs231775 with AR was assessed during the first year post-transplant. Other study characteristics, including immunosuppressive drugs, criteria used for acute rejection diagnosis, and the genotyping method of CTLA-4 rs231775 are reported in Supplementary Material, Table 46. Individual scores for each item of NOS in the identified studies are shown in Supplementary Material, Table CTLA-4 rs231775 was found in Hardy–Weinberg equilibrium (HWE) in 9 studiesA total of 15 cohorts including 5,401 KTRs were available for the meta-analysis of CTLA-4 rs231775 under the recessive contrast model (GG vs. AG/AA), while 13 cohorts enrolling 4,443 KTRs were available for the allelic (G vs. A) or the dominant (GG/AG vs. AA) model. A summary of random meta-analyses on the effect of CTLA-4 rs231775 on the risk of acute renal graft rejection is shown in Table In overall analyses, no evidence of publication bias or small study effects was found for the GG/AG vs. AA model  and dominant  models (Table 2 = 11%). In order to explore possible reasons for the observed heterogeneity, we conducted a subgroup analysis stratified by ethnicity, as well as three sensitivity analyses by using the following inclusion criteria: higher study quality (NOS ≥ 7), conformation with HWE, and AR within one year after kidney transplantation. A significant association was detected among Asians for the allelic  or the recessive model  in absence of between-study heterogeneity (I2 = 0%), while no association with CTLA-4 rs231775 was found among Caucasian or African patients. Results of sensitivity analyses provided evidence of a higher risk of acute rejection in the first year after kidney transplantation among carriers of the rs231775GG genotype , while no associations were detected among studies conforming to HWE or with a higher quality score (NOS ≥ 7)  of subgroup and sensitivity analyses were further investigated by using the FPRP test. At the pre-specified prior probability level of 0.001 to detect ORs of 1.50 , the FPRP values for the association of rs231775G or rs231775GG among Asians were 0.879 and 0.909, respectively, while the FPRP value for the association of rs231775GG with AR in the first year after kidney transplantation was 0.985 Table .Table 4S25, a higher risk of acute rejection has been reported in kidney transplant recipients under the GG vs. AA/AG model of CTLA-4 rs231775, but only the larger study25 also detected a higher risk under the G vs. A model contrast. Given the recent publication of two novel primary studies on the risk of acute kidney transplant rejection30, we conducted an updated meta-analysis with trial sequential analysis (TSA) to better estimate the impact of CTLA-4 rs231775 and to determine whether the currently available evidence was sufficient and conclusive. In addition, false-positive report probability (FPRP) analysis was conducted to examine whether the significant findings of the present or previous meta-analyses were noteworthy. To our knowledge, Dong and colleagues, in 2008, first applied FPRP to assess noteworthiness of meta-analytic estimates in a field synopsis47. Since then, more than 40 meta-analyses of genetic association studies have been published in which the FPRP method was applied to examine noteworthiness of significant pooled estimates.In the two most recent meta-analyses25 also revealed a lack of noteworthy results in the relationship between CTLA-4 rs231775 and acute renal transplant rejection. Overall, the current findings confute results of the most recent and larger meta-analysis25, which however included only 2,581 kidney transplant recipients.This updated systematic review and meta-analysis, which included a total of 5,401 kidney transplant recipients, showed that CTLA-4 rs231775 is not a genetic determinant of acute rejection. Results from the traditional pooled analysis technique are corroborated by TSA, which provided conclusive evidence against a clinically relevant impact of CTLA-4 rs231775 on the risk of acute renal graft rejection under the allelic (G vs A), dominant (GG/AG vs AA) or the recessive model (GG vs. AG/AA). Furthermore, none of the positive findings detected in the subsequent subgroup and sensitivity analyses, such as association of the G allele or of the GG genotype with a higher risk in Asian KTRs, were found to be noteworthy by FPRP. Similarly, the application of the FPRP test to statistically significant results of previously published meta-analyses48 and cancer49. Results from traditional meta-analyses show that the GG genotype of rs231775, which is associated to lower CTLA-4 expression and hence to a higher T cell activation and proliferation, may confer susceptibility to development of autoimmune diseases, such as rheumatoid arthritis50, Hashimoto’s thyroiditis51 and myasthenia gravis52. On the other hand, individuals with higher expression of membrane CTLA-4, due to the rs231775 A allele, may be at risk of developing multiple types of cancer53. These and other evidence support the hypothesis that only an optimal CTLA-4 expression can ensure a state of self-tolerance54, being rs231775 G  and rs231775 A  of CTLA-4 associated, respectively, with susceptibility to autoimmunity and cancer. However, the relevance of CTLA-4 rs231775 has not been consistently reported in every disease condition and is even less pronounced in cancer than in autoimmune diseases54. Interestingly, a meta-analysis with TSA have provided convincing evidence for association of CTLA-4 rs231775 with Hashimoto's thyroiditis55, nevertheless further investigation is still needed to clarify the role of CTLA-4 rs231775 in different autoimmune disorders and cancer types. In this regard, future application of TSA and/or FPRP to updated meta-analyses could be of value for conclusive demonstration of CTLA-4 rs231775 as susceptibility risk factor for autoimmunity and cancer.As a key regulator of the immune response magnitude, CTLA-4 genetic variation has been placed at the center of attention by investigators also for a possible role in autoimmunity57, reporting no evidence of association between CTLA-4 rs231775 and acute renal graft rejection, were unavailable to provide genotypes for the rs231775 SNP. Therefore, our pooled estimates must be interpreted with caution given the lack of inclusion of all available studies in the meta-analysis. Nevertheless, our TSA results showed that sufficient cumulative evidence has been reached to conclusively exclude a clinically relevant association of the G allele or GG genotype of CTLA-4 rs231775 with a higher risk of acute renal graft rejection. Second, it should be noted that a funnel plot asymmetry, indicating potential publication bias, was indeed detected for the allelic model (G vs. A), which however included only 13 cohorts compared to the 15 cohorts available for the recessive model (GG vs. AG/AA). In spite of this, the statistical correction for this bias, by using the trim and fill method, confirmed the combined risk estimate towards a null effect. Third, the majority of KTRs comprised in the present meta-analysis was of Caucasian ancestry , therefore a clinically relevant impact of CTLA-4 rs231775 in other ethnic groups cannot be formally excluded. Fourth, given the large heterogeneity among studies in terms of immunosuppressive therapy, we were unable to evaluate the impact of rs231775 on the risk of acute renal rejection according to a specific immunosuppressive drug. Fifth, we cannot exclude that CTLA-4 polymorphic variants other than rs231775 might play a role as genetic predictive factors for acute renal rejection. In this regard, it should be noted that a number of meta-analyses have been published on the association between other polymorphisms of CTLA-4 and acute renal rejection59, however none of these studies applied TSA or FPRP for convincing evidence of an association between these additional CTLA-4 SNPs and acute renal allograft rejection. Finally, the lack of informative data in the identified studies precluded the possibility to adjust ORs for clinical confounding factors, and to investigate interaction effects of rs231775 with other polymorphic gene variants of CTLA-4.Despite strengths of the present meta-analysis, such as the use of TSA and FPRP, our findings should be interpreted in the light of the following limitations and considerations. First, we attempted to conduct a comprehensive systematic review to identify all potential relevant articles, nevertheless corresponding authors of some eligible publications were unable, or unavailable, to provide genotype distribution of CTLA-4 rs231775. Among these, corresponding authors of two genome-wide association studies (GWASs)In summary, findings of this systematic review and meta-analysis exclude a role of CTLA-4 rs231775 as a genetic risk factor for acute renal transplant rejection. This conclusion is strengthened by results of TSA, which provided conclusive evidence against a clinically relevant impact of CTLA-4 rs231775 on the risk of acute renal graft rejection in overall populations. Nevertheless, large studies comprising Asian or African kidney transplant recipients are still required to clarify the impact of CTLA-4 rs231775 on acute rejection risk in non-Caucasian populations. In addition, investigation is also warranted in kidney transplant recipients to evaluate whether CTLA-4 rs231775 may have an effect on the risk of acute rejection when analyzed in combination with other CTLA-4 gene variants.Supplementary file1 (PDF 112 kb)


The pooled result showed a significant association between ESR2 rs1256031 gene polymorphism and ocular disease .Studies that were selected based on inclusion criteria reported 5 and 4 single-nucleotide polymorphisms (SNPs) identified in the ESR2 rs1256031 gene polymorphism had a protective effect against ocular disease, which supports the hypothesis that the estrogen-signaling pathway through ERβ plays a pivotal role in the pathogenesis of ophthalmic disorders.The recessive genotype of  It has been reported that E2 exerts a neuroprotective effect in the retina and optic nerve.11 Moreover, E2 seems to be involved in several eye pathologies such as glaucoma, myopia, age-related maculopathy (ARM), and cataract.15 Furthermore, the prevalence of dry eye disease is predominantly found in females, particularly in the menopausal and postmenopausal age group.16 Together, these support the notion that estrogen is a contributing factor to the development of ocular disease.In mammals, expression of 17 Estrogen-mediated effects are mainly modulated by ERα and ERβ, which are encoded by the ESR1 and ESR2 genes, respectively.18 The association between genetic polymorphisms of ER genes and the risk of disease has been reported extensively, particularly in cancer.20 Two common SNPs have been identified in ESR1  and ESR2 .22 Thus, SNPs in ESR genes may directly or indirectly affect normal physiological functions of estrogen and may affect the risk of ocular disease. Therefore, in the present study, a meta-analysis of all eligible studies was performed to provide an accurate estimation of the association between ER polymorphisms and the risk of ocular disease.DNA polymorphisms are different DNA sequences that commonly occur among individuals and populations. The most common type of DNA polymorphism is the single nucleotide polymorphism (SNP), which is characterized by a single nucleotide change in the DNA sequence. Interestingly, however, alterations in DNA sequences may be directly or indirectly correlated with the development of disease through the modification of its expression and functional effect.ER gene polymorphisms and the risk of ocular disease and (2) case-control design. The genotypic frequency for the ER polymorphisms was tested by Hardy-Weinberg equilibrium (HWE). Associations between the ER polymorphisms and risk of ocular disease were estimated by calculating pooled odds ratios (OR) and 95% confidence interval (CI). The random-effect model was used to allow heterogeneity. Heterogeneity was evaluated with Q-test and I2. Egger’s regression test was used to assess publication bias. P value of <0.05 was indicative of statistical significance.A literature search was conducted from PubMed, Google Scholar, Scopus, and Web of Science. Keywords such as ER, polymorphisms, and ocular disease were used in combination. The literature search was updated until September 2019. The selection criteria were as follows: (1) evaluating the associations between 12 recruited 425 glaucoma patients  with the average age of 63.55, while Kosior-Jarecka et al.21 enrolled 235 glaucoma patients  with an average age of 75.7. Seitzman et al.14 and Imbert et al.23 recruited only female participants with an average age of 65 and 62.2, respectively. A study by Škiljić et al.15 did not provide information regarding the gender and age of the participants. However, the study evaluated estrogen-related polymorphism in age-related cataracts. All of the studies used age- and gender-matched controls. Thus, this current study mainly consisted of menopausal female participants.From the literature search, 5 studies were selected based on inclusion criteria. The characteristics of the selected studies are shown in ESR1  and ESR2  genes, respectively. All SNPs complied with the HWE (p>0.05). Finally, only 3 studies were included in our meta-analysis of the association between ESR1 (rs2234093 and rs9340799) and ESR2 (rs1256031 and rs4986938) polymorphisms with ocular disease. The pooled results on the association between ER polymorphisms and the risk of ocular disease are shown in ESR1 gene polymorphisms and risk of ocular disease. However, the recessive model of ESR2 rs1256031 gene polymorphism showed a 45% decrease in odds ratio , suggesting a protective effect of recessive genotype in ESR2 rs1256031 against ocular disease. No publication bias was observed for the association of ER polymorphisms and risk of ocular disease (PEgger test>0.05).Five and 4 SNPs occurred in the 24 Moreover, menopausal women treated with hormone replacement therapy show a reduction of intraocular pressure26, which suggests a protective effect of estrogen. However, there is a conflicting result in regards to cyp19a1 polymorphism. A woman with cyp19a1 (rs10046) polymorphism has a higher susceptibility to myopia24, while there is no evidence of cyp19a1 polymorphism being associated with the risk of cataract.15 Furthermore, Nishikawa et al.27 provided evidence that there is no change in sex steroid hormone levels in patients with vitreoretinal disease. Nonetheless, the role of estrogen on ocular disease needs further investigation.In this study, 4 ocular diseases  were included in the analysis. Interestingly, such cases are mostly correlated with increased age. Indeed, low estrogen levels are observed in menopausal women and have been associated with increased cytokine production and ocular disease.ESR2 rs1256031 gene polymorphism was correlated with a reduction of ocular disease risk, which has been also reported as a protective factor in developing type 2 diabetes mellitus.28 Mice lacking ERβ are more susceptible to in vivo injury to RPE cells29, which supports a protective effect of ERβ. It has been reported that the expression of ERβ is more abundant than ERα in the central nervous system30, including the retina. Moreover, the expression of ERβ is relatively more constant than ERα in the human eye31, suggesting a prominent role of ERβ in regulating normal physiological function in the eye.Because the action of estrogen depends on the interaction with its receptors, understanding ER genetic variations become important to evaluate the association of ER polymorphisms and the risk of ocular disease. This report indicated that the recessive model of There are limitations to this study. First, the number of studies included in the meta-analysis was relatively small due to the limited availability of published papers. From 5 published papers, only 3 were suitable for further analysis, while the rest evaluated different SNPs. Second, major and minor alleles might have different roles in the risk of ocular disease, resulting in heterogeneity of the studies. Third, although this study was mainly generated from the pooled data of menopausal women, some studies enrolled both male and female participants, which may affect the pooled estimate. Thus, these findings should be interpreted with caution.ESR2 rs1256031 gene polymorphism is significantly associated with the risk of ocular disease. The recessive genotype of ESR2 rs1256031 gene polymorphism was associated with a reduced risk of developing ocular disease. It is expected that more studies will become available, which may help the accurate estimation of the relationship of ER with ocular disease to verify this conclusion.In summary, the current meta-analysis suggests that


Background:miR-146a has been demonstrated to be involved in normal hematopoiesis and the pathogenesis of many hematological malignancies by inhibiting the expression of its targets. Rs2910164(G>C) may modify the expression of the miR-146a gene, which might influence an individual's predisposition to childhood acute lymphoblastic leukemia (ALL). However, inconsistent findings have been reported on the association between the rs2910164(G>C) polymorphism and the risk of childhood ALL.Methods: A comprehensive meta-analysis was performed to accurately estimate the association between the miR-146a rs2910164 polymorphism and childhood ALL among four different genetic models.Results: This meta-analysis included Asian studies with a total of 1,543 patients and 1,816 controls. We observed a significant difference between patients and controls for the additive model  using a random effects model. Meanwhile, there was a trend of increased childhood ALL risk in the dominant model , recessive model  and allele model  between patients and controls.Conclusions: Our findings suggest that the miR-146a rs2910164 CC genotype was significantly associated with childhood ALL susceptibility. Acute lymphoblastic leukemia (ALL) is a very common malignancy among children, accounting for ~75% of leukemia cases among children  signaling has been reported to be constitutively activated in a variety of leukemic cells. miR-146a could be upregulated by NF-κB  guidelines. The PRISMA checklist is provided in miR-146a or microRNA-146a” “rs2910164” “leukemia” with no language or time restrictions. All studies were assessed by reading the title and abstract and irrelevant studies were excluded. Then, the full texts of the remaining studies were assessed to determine their eligibility.The PubMed, Google Scholar, WanFang and Chinese National Knowledge Infrastructure databases were systematically searched for relevant studies using the words “miR-146a rs2910164 polymorphism and the risk of childhood ALL; (b) studies in which all patients had been diagnosed with ALL by morphology, immunology, cytogenetic and molecular biology (MICM) in accordance with one of the following criteria: (i) bone marrow morphology standard: according to the 2016 WHO diagnostic criteria, the original and immature lymphocytes in bone marrow are not <20%, (ii) if the original and immature lymphocytes were <20%, a molecular diagnosis method was used to determine whether ALL pathogenic genes existed; (c) studies with data that could be used to estimate odds ratios (ORs) with corresponding 95% confidence intervals (CIs); and (d) studies published before April 30, 2020.The study inclusion criteria were as follows: (a) case-control or cohort studies that assessed the association between the The exclusion criteria were as follows: (a) the study was not a case–control study; (b) the study was not related to acute lymphoblastic leukemia or 146a rs2910164; (c) the study lacked particular genotype data; (d) the genotype distribution of the control subjects was not in a Hardy–Weinberg equilibrium (HWE).The following data were independently extracted from included studies and entered into a database to ensure the veracity of the data: first author's name, year of publication, population, genotyping techniques, number of patients and controls, genotype distribution, allele distribution, HWE and other information. Studies were excluded if they did not provide the above information.I2-test. I2 statistics range from 0 to 100%. Significant heterogeneity was defined as P < 0.01 and I2 > 50%. ORs with corresponding 95% CIs were calculated using the fixed effects model (Mantel-Haenszel) when no significant heterogeneity was observed; otherwise, a random effects model was used. The Z-test was used to test the significance of the ORs. To check the stability of our results, sensitivity analyses for the overall effect were conducted by excluding one study at a time. Additionally, Egger's and Begg's tests were used to assess publication bias. The statistical analyses were performed using the STATA v.16.0 software , Review Manager 5.0.24 , R version 3.6.2  and RStudio version 1.2.1 .The Hardy-Weinberg equilibrium was examined by Pearson's chi-squared test. Four genetic models were used in the study: the dominant model (CC +CG vs. GG), the recessive model (CC vs. GG + CG), the additive model (CC vs. GG) and the allele model (C vs. G). Genetic heterogeneity was evaluated using the Q-test and A total of 113 potential studies were retrieved through the initial search. Twenty duplicates were excluded. Then, the titles and abstracts of 93 studies were screened and 73 studies were excluded. The full texts of the remaining 20 articles were evaluated; 6 were excluded because they were not case-control studies and 7 were excluded because they were not related to acute lymphoblastic leukemia or rs2910164 and 1 study was excluded because it did not provide sufficient data. A flow chart of the study selection process is shown in I2 statistics shown in I2 = 74.9%), CC vs. GG + CG (I2 = 69.9%), CC vs. GG (I2 = 81.6%) and C vs. G (I2 = 79.5%) models for the rs2910164 polymorphism. As high heterogeneity was observed, sensitivity analysis was performed to analyze the sources of heterogeneity.The Cochran's Q test and the I2 = 0.0%, P = 0.571). The other three models showed a moderate degree of heterogeneity   .P = 0.049) and a trend of increased childhood ALL risk for the dominant model , recessive model  and allele model  between patients and controls  and the funnel plot was symmetrical, as the studies did not coagulate into one quadrant of the funnel  ranged from 0.395 to 0.495 in each study  can be found in the article/DZ, JY, ZY, and QZ were responsible for the statistical analysis, study design, and manuscript preparation. DZ and CT managed the literature searches and analyses. This study was supervised by YW, QC, and RC. 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.


TNF polymorphism (rs1799964) modifies this association. Semi-quantitative food frequency questionnaire derived data were used to calculate the DII scores. Odds ratios (OR) and 95% confidence intervals (CI) were calculated using multivariable logistic models adjusted for confounders. When we stratified the data by sex, the association between GC and the DII was significant only among the women , and the DII effect on the risk of GC differed depending on the TNF genotype . Furthermore, the association between the DII and GC was significant in the Helicobacter pylori-positive group; similarly, the effect differed based on the TNF genotype . In conclusion, rs1799964 may modify the effect of the DII on GC.The inflammatory process is known to increase the risk of gastric carcinogenesis, and both genetic and dietary factors are associated with inflammation. In the present study of 1,125 participants (373 cases and 752 controls), we determined whether the dietary inflammatory index (DII) is associated with the risk of gastric cancer (GC) and investigated whether a  In 2016, the incidence of GC was first and second among all types of cancer in Korea2. Such a high prevalence of GC may be attributed to Helicobacter pylori infection and chronic inflammation caused by chronic gastritis3.According to the GLOBOCAN results reported in 2018, gastric cancer (GC) is the fifth most common cancer worldwide, and the highest mortality rates are found in East Asia, including Korea5. Among these cytokines, tumor necrosis factor (TNF) is among the most important factors inhibiting gastric acid secretion, causing the development of GC6. TNF production is generally regulated at the transcriptional level6. Polymorphisms located in the promoter region of the TNF gene affect the level of TNF. Indeed, genetic polymorphisms of the TNF gene have been proposed as candidate risk factors for GC7. However, the results of studies investigating the association between TNF genetic polymorphism and GC risk have been inconsistent9. These results indicate the need for further studies considering other factors, such as individual lifestyle factors.The inflammatory process is caused by cytokine secretion and increases the risk of the development of gastric carcinogenesis10. For example, the consumption of fruits, vegetables, and beans containing antioxidants has been reported to help reduce inflammation, whereas the consumption of saturated fats, refined carbohydrates, and processed meat can cause inflammation12. However, the analysis of a single food group or single nutrient does not cover the entire intake of an individual, and there is a limit to identifying the association with the disease13. Since the results of most studies focusing on single nutrients are inconsistent, it is necessary to assess the overall effect of diet-induced inflammation reflecting the combination of nutrients consumed by individuals on the risk of cancer15. Moreover, the genetic sensitivity of the host can control the effects of the disease through complex interactions with environmental factors, and diets containing anticancer nutrients and carcinogens can modulate the risk of cancer development, especially in genetically vulnerable individuals17. Thus, gene-diet interactions can account for various outcome differences in individual GC risk, complementing the findings of studies investigating a single gene or a single diet.Environmental factors and host genetic factors also contribute to the chronic inflammatory response. Among these factors, dietary components contain both carcinogens and anticancer substances and are known to control the risk of cancerTNF genetic polymorphism is related to the risk of GC. In particular, we focus on identifying the association between GC and a genetic polymorphism of TNF (rs1799964); the association between rs1799964 and both GC and H. pylori infection has been previously reported19. We aimed to study whether differences in individual TNF genes have different effects on GC depending on eating habits and vice versa.In this study, we found that dietary inflammation is associated with the risk of GC. We also analyzed whether a H. pylori infection was higher among the cases (84.5%) than the controls (40.4%). These variables might be associated with the GC risk and, therefore, were included in the subsequent statistical analyses that considered potential confounders. The total energy intake and DII score of the cases were significantly higher than those of the controls (p < 0.001). However, no differences were found in the distribution of a family history of GC or alcohol consumption between the cases and controls. Additionally, regarding age and sex, there was no difference because the participants were matched beforehand.Table H. pylori infection status along with the variables included in Model 1. In the crude model, the participants in the highest DII group had a 1.70-fold higher GC risk than those in the lowest DII group (odds ratios (OR), 1.70; 95% confidence intervals (CI) 1.24–2.33). This association was also significant after adjusting for all confounding factors . When stratified by sex, compared to the risk associated with the lowest DII score group, the women with the highest DII score showed an association with an increased risk of GC, but there was no significant association among the men . The women also showed a trend of significantly increased GC risk as the DII increased . Additionally, sex showed a significant interaction with the DII in the context of the GC risk (p for interaction = 0.014). Stratified by the H. pylori infection status, the association between the DII and GC was observed only in the H. pylori-positive group , but the interaction between the DII and infection status was not significant (p = 0.36) (Supplementary Table Table TNF (rs1799964) and the GC risk are shown in Table The results of the association between a genetic polymorphism of TNF gene modifies the association between the DII and GC risk, we performed a stratified analysis by individual genotype . However, with C, there was no significant association between the DII and GC risk, although the interaction effect was suggestive (p for interaction = 0.055). A significant interaction was observed among the women. Among the women, the TT homozygotes of rs1799964 in the high DII group exhibited an increased GC risk by 2.30 times compared to the risk in the low DII group . However, among the C carriers, there was no statistical association between the DII and GC, although there was a decrease in the GC risk in the high DII groups . When we further adjusted for the H. pylori infection status in Model 2, a significant association between high DII scores and the GC risk was also observed only in the TT genotype carriers, although the statistical significance of the interaction disappeared (p for interaction = 0.068). In the case of TT genotype, an inflammatory diet increased the risk of GC by nearly more than 2 times compared with the risk of the C carrier group when comparing the high and low DII groups . Among the men, a significant interaction between the DII and TNF gene polymorphism could not be identified . Stratified by the H. pylori infection status, the same results were observed in the H. pylori-positive group. The TT homozygotes showed an increased risk of GC in the high DII group . However, among the individuals with the C allele, the risk of GC in the high DII group was decreased, although the association with DII and GC was not significant . But, no association with DII and GC or an interaction effect was observed in the H. pylori-negative group , we observed a significant association between the DII and risk of GC in women but not men. We also identified significant modification effects on the association between the DII and risk of GC according to a 20. Although many GC studies focus on H. pylori infection, environment-induced inflammatory reactions or genetic differences can also increase the risk of GC. Among such environmental factors, we focused on diet factors considered to induce inflammation through oxidative stress reactions, and we could confirm that an inflammatory-inducing eating habit is more dangerous in women, which is consistent with previous research results22. These results may be attributed to differences in eating habits between men and women and the effects of sex hormones. According to previous studies investigating sex differences in the immune response, inflammation-related genes can be overexpressed in women due to differences in hormone conditions and rare genes on X chromosomes, and this overexpression results in poor prognoses among women suffering from chronic inflammatory diseases24. Additionally, in the present study, the men and women had different drinking habits. In the case of the women, 43% were current alcohol consumers, whereas 73% of the men were current alcohol consumers  secreted by immune cells, leading to changes in the epigenomeTNF have been shown to contribute to differences in individual inflammatory reactions25. Sugimoto et al.19 reported that rs1799964 C alleles were involved in the susceptibility to GC and can induce higher TNF production, especially in East Asian individuals. Additionally, rs1799964 T>C polymorphisms might be associated with a reduced risk of H. pylori infection18. Therefore, the association with GC risk in that region was expected, and a previous meta-analysis suggested the need for further studies to provide evidence supporting the association between TNF-1031 (rs1799964) and GC9, but such an association was not confirmed in our study. However, interestingly, according to our results, the effects of an inflammatory diet on the GC risk differ depending on the TNF rs1799964 genotype. The exact mechanism of this phenomenon has not been explained, but it is possible that the variation in the promoter region is related to gene expression6.Another factor related to the inflammatory response is the genetic predisposition of the host. In particular, single nucleotide polymorphisms (SNPs) present in the promoter of TNF-α is a multifunctional cytokine produced by monocytes and macrophages that plays an important role in promoting the inflammatory response and cell proliferation and inducing apoptosis6. Polymorphisms located in the promoter region of the TNF gene affect the transcription of theses genes6. Some studies have reported a correlation between polymorphism rs1799964 and the TNF mRNA expression level27. Nourian et al.28 demonstrated that the level of TNF mRNA expression was reduced in individuals with the genotype TT compared with that in individuals with other genotypes. Considering the mechanism by which the substances contained in the diet induce the secretion of cytokines and cause the inflammatory cell response29, we assume that if cytokine secretion is genetically altered, it can be difficult to identify the same influence across different genetic types.H. pylori infection status, the results suggested that these risky diet patterns can cause different cancer development depending on the individual's genetic background. However, the interpretation requires caution, and studies involving larger samples are warranted because the number of cases in the H. pylori-negative group was very small. Additionally, other types of cytokines, such as the interleukin (IL) family , contribute to the inflammatory response31. Indeed, many studies have reported abnormal IL concentrations in GC patients and suggested promising indicators of GC development33. Therefore, further studies are needed to identify whether genetic polymorphisms associated with other cytokines can modify the diet effect that increases the risk of cancer.Our results suggest that inflammation-inducing eating habits can result in different outcomes depending on the individual's genetic background, and it is believed that dietary interventions considering individual genotypes could help prevent GC early on. In addition, when we conducted further stratification analysis by the TNF. Another strength of this study is that we calculated an index score to reflect the inflammatory potential of a diet, assessed its association with the risk of GC and used it to examine interactions with genes, which can consider the overall intake conditions and an individual’s genetic background rather than only one risk factor. However, our study has several limitations. First, because of the relatively small number of GC cases, we have insufficient statistical power. Second, although we had information regarding the cardiac or noncardiac cancer status of the GC cases, because the number of cases was small, we were unable to observe a difference in the GC risk according to cancer subtypes. Additionally, this study may have been affected by selection bias due to its hospital-based case–control study design.To the best of our knowledge, this study is the first to report a gene-diet interaction while particularly focusing on a gene functionally associated with secreted serum TNF polymorphism and GC, rs1799964 may modulate the effect of diet on GC, and a pro-inflammatory diet had a greater effect in the general population who did not have a genetic variant. Therefore, these results may support dietary recommendations for GC prevention and therapy on the basis of gene-environment interactions. To prevent GC, it seems important to reduce the amount of food-induced inflammation and consume a diet reportedly associated with anti-inflammation, such as fruits and vegetables. Further evaluations of the associations of the dietary factors included in the DII calculation and studies of other cytokine genetic polymorphisms are necessary to better understand the heterogeneity in the etiology of GC.In conclusion, our findings demonstrate that the inflammatory potential of diets is associated with an increase in the risk of GC, particularly among women. Although there was no evidence of an association between the 34. In brief, cases who were diagnosed with GC 3 months before recruitment were defined as patients. Cases diagnosed with other cancers within 5 years, diabetes mellitus, or severe mental disease and pregnant women were excluded. The controls were selected among individuals who visited the Center for Cancer Prevention and Detection at the NCC, and subjects with a history of cancer, diabetes mellitus, gastric ulcers, and H. pylori treatment were excluded. The cases and controls were matched for sex and 5-year age distributions. In total, 373 GC cases and 752 healthy controls were included in this study, and their genotype and dietary intake data were available  in Korea between March 2011 and December 2014. The details of this study have been previously described35. In brief, genomic DNA was extracted from peripheral blood, and the Affymetrix Axiom Exom 319 Array  platform, including 318,983 variants, was used for genotyping. Genotype imputation was performed using the Asian population (n = 504) in 1,000 Genome haplotypes phase III integrated variant set release GRch37/hg19 (https://www.1000genomes.org/) as a reference panel. Genetic markers with a minor allele frequency (MAF) < 0.01, call rate < 95%, and deviation from the Hardy–Weinberg equilibrium  were discarded. SHAPIT (v2.r837) and IMPUTE2 (2.3.2) were used to perform the phasing and SNP imputation, respectively. The quality control criteria were applied after filtering for an INFO score over 0.6. Finally, the genetic polymorphism (rs1799964) in the TNF promoter was selected.The genotyping and quality control process were presented in detail in a previous study36. The participants provided their individual average frequency of intake and portion sizes of specific foods in the year preceding the interview. These values were converted to obtain the daily nutrient intake using a scale with nine frequency categories  and three portion size categories  included in the SQFFQ. The total energy and nutrient intake of each participant were analyzed by the Computer Aided Nutritional analysis program . The SQFFQ-derived data were used to calculate the DII scores of all participants. The details of the development37 and construct validation39 of the DII have been previously described. The dietary data were used to calculate the z-score of each food parameter for each individual. The global means and standard deviations of the food and nutrient intakes collected from 11 nations, including Korea, were used to calculate the z-score. Then, the z-score was converted to a percentile and centered by doubling the value and subtracting 1. The centered percentile value was multiplied by the respective inflammatory effect score to obtain the food parameter-specific DII score37. The overall DII score was calculated as the sum of all available food parameter-specific DII scores. This study included the following 35 of the original 45 DII food parameters to generate the DII score: protein, fat, carbohydrate, fiber, monounsaturated fatty acid (MUFA), polyunsaturated fatty acid (PUFA), saturated fatty acid, n-3 fatty acids, n-6 fatty acids, cholesterol, thiamin, riboflavin, niacin, vitamin B6, vitamin B12, vitamin C, folic acid, vitamin A, vitamin D, vitamin E, β-carotene, iron, magnesium, selenium, zinc, garlic, ginger, onion, green tea flavan-3-ols, flavones, flavonols, flavanones, anthocyanidins, and isoflavones. Energy was not used to compute the DII because all components of the DII were adjusted for energy intake using the energy density approach, which was calculated per 1,000 kcal of energy40. The world database was also standardized to 1,000 kcal/day.The dietary intake of each participant was assessed using a 106-item semiquantitative food frequency questionnaire (SQFFQ). The validity and reproducibility of the SQFFQ have been previously reportedH. pylori infection status in Model 1. A stratified analysis by sex and H. pylori infection status was performed, and the median value of each DII tertile was used as a continuous variable to test for trends. Genetic association analyses were performed in the dominant models. To analyze the interaction effects between the DII score and the genetic polymorphism, low- and high-score groups were categorized according to the median DII score of the control group. The interaction between the DII and SNPs was analyzed by using the likelihood ratio test between the models with and without the interaction term (DII*SNPs). All statistical analyses were performed using PLINK version 1.0741 and R software version 3.4.3.The difference in general characteristics between the cases and control subjects were analyzed by using the chi-square test  or t-test (continuous variables). The DII score was categorized into tertiles based on the control’s score to investigate the association between inflammatory intake and the GC risk. The OR and 95% CI of the GC risk were calculated for each group using a univariate logistic regression for the crude model and a multiple logistic regression with the lowest tertile as a reference group. The multivariable model (Model 1) was adjusted for age, sex, education level, smoking status, and physical activity. Model 2 additionally adjusted for the Supplementary Information 1.


Dear Editor,1 published in the European Journal of Internal Medicine and entitled ‘Active smoking is not associated with severity of coronavirus disease 2019 (COVID-19)’ had errors and led to the wrong conclusion.The letter to the Editor of Lippi and Henry1 searched PubMed and Web of Science up to 9 March 2020, and identified 5 studies2 with data on smoking and severity of COVID-19. They performed a meta-analysis revealing a pooled OR of 1.69 (95% CI: 0.41–6.92) and concluded that active smoking does not seem to be significantly associated with enhanced risk of progressing towards severe disease in COVID-19. There were several mistakes in their data collection that led to errors in the meta-analysis. In table 1 of their letter, they indicated the outcome of Guan et al. study2 to be ‘Admission to ICU, mechanical ventilation, death’, however, they used the data of ‘severe disease’ in the study. According to the Guan et al.2 paper, the number of patients having composite outcome should be 66, and for patients not having a composite outcome should be 1019. However, Lippi and Henry1 used 172 and 913, respectively, in their paper. This is the most serious mistake because the Guan et al.1 study contributes to most of the cases in the meta-analysis. Moreover, the non-severe patients in the Huang et al.3 study should be 28 and not 31. The non-severe patients in the Yang et al.5 study should be 20 and not 18. The severe patients in the Zhang et al.6 should be 58 and not 60. The errors led to the wrong sample size of these 3 studies as well. Lippi and Henry1 were only correct in one4 out of the 5 studies.Lippi and Henry1 paper, the purpose of this letter is to correct their errors, therefore new studies are not included in the updated meta-analysis.I performed an updated meta-analysis according to the correct data using RevMan Ver. 5.3, and provide the forest plot . The poo7, 5 studies were included. Though meta-analysis was not performed in that study, the authors concluded that smoking is most likely associated with the negative progression and adverse outcomes of COVID-19. A recent meta-analysis including 7 studies also revealed that smokers have a double risk of severe COVID-19 8. A meta-analysis published in 2019 including 27 studies and 460592 participants revealed current smokers  and ex-smokers  were more likely to develop community-acquired pneumonia compared to never smokers9. The evidence suggests that smokers are more vulnerable to lung infection, and COVID-19 is no exception.In a systemic review published by Vardavas and Nikitara1 paper should have had different results.In conclusion, the results of this updated meta-analysis suggest that active smoking is significantly associated with the risk of severe COVID-19. Though more data are available now, they are not included in this study. However, the early meta-analysis of the Lippi and Henry


Many contradictory conclusions pertaining to the telomere length in peripheral leukocyte chromosomes as a potential biomarker for ischemic stroke (IS) risk have been reported by the various observational studies in previous years. This study aims to investigate whether the leukocyte telomere length is associated with an increased IS risk or not, based on the Mendelian randomization (MR) approach.Q test, and the fractional polynomial test).Based on the NHGRI-EBI GWAS Catalog database, the Chinese online genetic database as well as the previous published studies, twelve single nucleotide polymorphisms (SNPs) with minor allele frequency ≥ 0.05 were selected and the leukocyte telomere length was measured in 431 first-ever IS patients and 304 healthy controls (quantitative polymerase chain reaction). To explore linear and non-linear effect of telomere length on the IS risk, we preformed the linear MR analysis , and the non-linear MR analysis  were chosen as instrumental variables. In linear MR analysis, the adjusted odds ratios and 95% confidence intervals of IS for genetically predicted telomere lengths, based on the two SNPs, were 1.312 (0.979 to 1.759), 1.326 (0.932 to 1.888) and 1.226 (0.844 to 1.781) for the inverse-variance weighted method, the maximum likelihood method, and the mode-based estimation method, respectively. Three tests for nonlinearity failed to reject the null exactly, indicating that the relationship between telomere length and IS risk is unlikely to be non-linear.This MR study based on individual data does not provide strong evidence for a positive linear or non-linear effect of telomere length on the IS risk. In China, stroke is one of the leading causes of death, and about 78% of all strokes are ischemic stroke (IS) . IS is aConventional observational studies have shown inconclusive associations of telomere shortening and the risk of stroke. Consistently, several original studies reported positive association between telomere shortening and the risk of stroke –14. In cRandomized controlled trials of interventions specific to TL have not yet been done in relation to stroke outcomes. In the absence of such trials, Mendelian randomization (MR) analysis focused on genetic instrument can be used to investigate the causality from routinely observational studies . Based oAll participants were recruited from Jidong Oil-field Hospital, Chinese National Petroleum, and Beijing Tiantan Hospital, Capital Medical University, during 2010–2013. A total of 755 participants aged 18 or above were found to be eligible. All participants who met any of the following criteria were excluded from the study: (1) history of mental illness or infectious disease; (2) history of aneurysm caused by cerebral infarction, cerebral haemorrhage or other cerebrovascular diseases, congenital heart disease, acute myocardial infarction, liver disease, renal failure, malignant tumour, chronic obstructive pulmonary disease, rheumatoid arthritis, or other diseases; and (3) pregnant or lactating women. All first‐ever IS patients were diagnosed according to the World Health Organization criteria . In thisThis study was approved by the Ethics Committee of Capital Medical University, China (No. 2016SY23). This study was in accordance with the principles of the Declaration of Helsinki. All participants provided their written informed consent before taking part in this study.Blood samples were collected and processed according to the standardized protocol. Following 10 h. of overnight fasting, blood samples were collected by venipuncture in two different tubes containing an anticoagulant and a coagulant respectively. Samples were processed within 8 h. and stored at -80C until further measurement. Given that whole blood TL is a proxy for tissue-specific TL for many tissues, and blood TL is a proxy for TL in some tissues in epidemiological studies , in this36B4) which encodes the acidic ribosomal phosphoprotein P0, was used to normalize the quantity of the input DNA. Comparing telomere-to-36B4 ratio to reference DNA of known TL reflects the absolute length of the telomeres. The master mix was prepared using Applied Biosystems reagents. Primer concentrations were: TelF 200 nM, TelR 200 nM, 36B4F 300 nM, and 36B4R 500 nM. The primer sequences (5′–3′) were as follows:Serum samples were used to measure the biochemical indices, and blood cells were used to extract the genomic DNA. Based on automated nucleic acid purification platform , genomic DNA was extracted from the 200 μl of frozen blood cells using the magnetic bead-based method for concentrating DNA . DNA concentration and purity were determined using a Nanodrop  and then normalized to 5 ng/μl in TE buffer . Absolute TL of each chromosome end of the peripheral leukocytes was measured by using a validated quantitative polymerase chain reaction (qPCR) on the method reported by O'Callaghan with modification , 29. A sTelF CGGTTTGTTTGGGTTTGGGTTTGGGTTTGGGTTTGGGTTTelR GGCTTGCCTTACCCTTACCCTTACCCTTACCCTTACCCT36B4F CAGCAAGTGGGAAGGTGTAATCC36B4R CCCATTCTATCATCAACGGGTACAAAnalysis of both telomere and 36B4 in experimental samples and reference DNA were run in triplicates using 5 ng DNA and both telomere and 36B4 amplifications were performed on the same run. A no template control of nuclease-free water was included in each run. In brief, qPCR was performed in 0.1 ml tubes on the Applied Biosystems QuantStudio 3 Real-Time PCR System , with the thermal cycling profile for both telomere and 36B4 amplifications: 2 min at 50 °C, 2 min at 95 °C, followed by 35 cycles of 95 °C for 15 s, 61 °C for 1 min, followed by a melt curve. The no template controls in all runs were no amplification. The melt curve should produce only one single distinct peak. Standard deviation for the Ct-value in replicates were less than 0.5. The inter-coefficients of variance for the Ct-value was less than 5%.https://www.ebi.ac.uk/gwas) and the online genetic database (https://www.ensembl.org) according to reported associations with TL and minor allele frequency (MAF) ≥ 0.05. Only one SNP (rs17653722) is reported in population of Chinese descent and other TL-related SNPs were based on the European ancestry. The exclusion of association could be confirmed in online software PhenoScanner (https://www.phenoscanner.medschl.cam.ac.uk/phenoscanner). Finally, twelve SNPs  were selected for analysis. SNP genotyping was performed on these SNPs using the Sequenom Massarray iPLEX platform .Single nucleotide polymorphisms (SNPs) were selected from the NHGRI-EBI GWAS Catalog , systolic blood pressure (SBP), diastolic blood pressure (DBP), were collected by questionnaire survey and physical examination. Biochemical data including fasting plasma glucose (FPG), triglycerides (TG), total cholesterol (TC), high‐density lipoprotein cholesterol (HDL‐C), low‐density lipoprotein cholesterol (LDL‐C), apolipoprotein A1 (ApoA1), apolipoprotein B (ApoB) were all tested via standard methods in the clinical laboratory of Beijing Tiantan Hospital. TL measurement and SNP genotyping were performed by laboratory personnel according to the standardized protocol.t test or the Mann–Whitney U test. The chi‐squared test was used to compare the proportions for categorical variables, and to test for Hardy–Weinberg equilibrium.Demographic and clinical characteristics were represented as the mean ± standard deviation (SD) for continuous variables underlying the normal distribution; otherwise, the median (interquartile range) was used. Categorical variables were represented as frequency (percentage). The between-group differences of continuous variables were tested by the The genetic variant or genetic risk score (GRS) acts as an instrumental variable in MR analysis if: (1) they are truly associated with TL; (2) they are independent of other factors (confounders); and (3) they can only influence IS risk via the causal effect of the TL. All association of instrument variables with TL and other risk factors, which were susceptible to reverse association, were limited to health controls. Principal analyses assumed over-dominant effects (heterosis), with subsidiary analyses of other genetic models .β coefficients were obtained from the linear regression model with natural log-transformed TL (ln TL). The percentage difference in TL with risk genotype was obtained from [exp (β) − 1)] × 100. Then, the linear MR estimates for ln TL on the IS risk were calculated by the inverse-variance weighted method, the maximum likelihood method, and the mode-based estimation method, adjusting for age, sex, and other confounders . All results were presented as the odds ratio (OR) of IS per 10% decrease in TL. Three tests for non-linearity of the semiparametric method  were applied: the quadratic test, which assesses for a linear trend among the localised average causal effect (LACE) estimates, Cochran’s Q test, which assesses whether LACE estimates differ more than would be expected by chance, and the fractional polynomial test, which assesses whether the fractional polynomial model of degree 1 fits LACE estimates better than the linear model [The ar model .P value < 0.05 was considered to be statistically significant. All statistical analyses were performed using R version 3.5.3 .For all analyses, a two‐tailed 25 44 years, P75 61 years) in all subjects, 47 years (35 to 59 years) in controls, and 57 years (49 to 63 years) in IS patients. Most of the IS patients were older males with smoking and drinking habits and higher weight, BMI, SBP, DBP, TG levels and lower TC, HDL‐C, ApoA1, ApoB levels. The levels of TL, height, FPG, LDL‐C, and ApoB/ApoA1 ratio were not statistically different between groups. There were no statistically significant differences in the genotype distribution of all SNP between the IS patients and the control group . The ORs for IS of risk genotype (“short TL”) were 0.810  for rs11125529, and 0.890 (0.660 to 1.190) for rs412658  for IS was 0.681 (0.469 to 0.982) adjusted for age and sex , indicating that the effect of telomere attrition on IS risk may not be non-linear. In Fig. Using rs11125529 and rs412658 as a proxy, the linear MR analysis provided no evidence of an overall association between genetically predicted TL and IS risk (OR (95% CI): 1.312 (0.979 to 1.759) for the inverse-variance weighted method, 1.326 (0.932 to 1.888) for the maximum likelihood method, and 1.226 (0.844 to 1.781) for the mode-based estimation method), after adjusting for the above-mentioned factors  as proxies to clarify whether TL is causally relevant to IS or not. Our results showed no causal association between the genetically shortened TL and the increased IS risk.P = 0.42) [P = 0.46) [P = 0.60) [Many epidemiologic studies have examined the relationship between TL and stroke risk, but results were not consistent enough. Shorter TL were not significantly associated with IS in the prospective and retrospective studies carried out in the Caucasian , 30, 31. = 0.42) . Consist = 0.60) . In anot = 0.60) . The preI2 = 81.1%, Phet < 0.001) [Despite previous retrospective observational studies reporting an TL-stroke association, such a relationship has not been firmly supported by evidence from prospective cohort studies. Subgroup analysis of four meta-analysis (stratified by study design) have shown consistent results that shorter TL were not significantly associated with stroke risk in prospective studies or in studies with a high quality score , 33, 34.< 0.001) . The incThree linear MR studies based on two-sample MR design also presented inconsistent results in European ancestry. One MR analysis showed conflicting results which shorter TL was marginally statistically significantly associated with the decreased risk of stroke . Anotherr2 = 0.75, P < 0.0001) with the results obtained with terminal restriction fragment analysis (the gold standard for TL measurement). However, slightly difference in TL measurement could affect the TL-IS relationship [There are several possible explanations for the discrepancy between retrospective and prospective cohort studies. One hypothesis is that TL had causal effect on IS risk under certain circumstances. Difference in factors, such as participant age range, TL measurement technique, and so on, might influence the TL-IS relationship. A previous study indicate that the positive association between short TL and the risk of stroke or post-stroke death might only exist in the seniors population (ranging from 65 to 73 years old), therefore the effect of age might need to be taken into consideration in future studies . Howevertionship .Another hypothesis, based on epidemiological evidence, explaining this contradiction is that shorter TL is inversely associated with the risk of IS, which means that TL may be a downstream biological consequence of the IS onset. Figure For the chief strengths of our study, we explored the possible shape of the potential causal relation between TL and IS risk in a one-sample MR framework using linear and non-linear MR methodology. As a result of the MR analysis, potential reverse causality was eliminated and confounding bias was reduced because genetic instruments were not associated with individual risk factors that may affect results from conventional observational studies. Secondly, although the potential pleiotropic effects were unavoidable in this study, we searched comprehensively from genotype to phenotype to identify the potential pleiotropic effects and further provided possible evidence of the SNP instrument validity that the SNPs have no effect on available confounding factors, to reduce the likelihood of bias due to violation of the instrumental variable analysis. Furthermore, to our knowledge, this is the first linear and non-linear MR study assessing TL in relation to the IS risk in a Han Chinese population. Otherwise, healthy controls were randomly recruited from the general population covering same geographical area, which could decrease the selection bias of the results.This study also has some limitations. Firstly, MR analysis has stringent assumptions . CompletIn conclusion, although TL is associated with the risk of IS based on the conventional case–control analysis, this one-sample MR study suggests that negative association of TL with the IS risk is unlikely to be linear or non-linear causal. There is a need to identify the specific genetic, biochemical, and environmental biological mechanisms responsible for this association. Further research with larger sample sizes, which will be able to perform stratified analysis by age or other strong risk factors, is necessary to understand the causal pathways underpinning this association.Additional file 1: Table S1. Association between SNP genotypes and telomere length under co-dominant and additive model.Additional file 1: Table S2. Association between SNP genotypes and telomere length under dominant, recessive and over-dominant model.


Background: Long intergenic non-protein coding RNA 511 (LINC00511) is upregulated in diverse cancers and involved in prognosis. This study aimed to evaluate the prognostic profile of LINC00511 in cancer patients.Methods: Published studies evaluating the prognosis of LINC00511 in patients with different cancers were identified from Medline, Embase, and Web of Science. Analysis of the association between LINC00511 and clinicopathological characteristics was conducted. GEPIA was used to validation and functional analysis and LnCeVar was used to get genomic variations.Results: We eventually included 9 studies, and the combined results showed LINC00511 was significantly associated with decreased OS  albeit with mild heterogeneity , similarly in cancer type subgroups: breast cancer, digestive system cancer, and cervical cancer . There is no publication bias and meta-regression indicated follow-up time maybe heterogeneity of the results (P = 0.008). Additionally, LINC00511 appeared to be correlated with age, clinical stage, tumor size, and lymph node metastasis. Those findings were confirmed in GEPIA. Through LnCeVars, gene ontology and functional pathways were enriched, and dysregulated hallmarks and related ceRNA network of LINC00511 were disturbed.Conclusions: LINC00511 could be predictive of poor OS and lymph node metastasis in multiple cancers, in another word, LINC00511 serves as an unfavorable prognostic factor, and its mechanism is related to ceRNA. Cancer is becoming a major threat to human health and a major public health problem worldwide with its increasing morbidity and mortality  are a kind of different RNAs, which has longer than 200 nt transcripts with disability of encoding proteins  searched Medline, Embase and Web of Science for cohort studies comparing the relationship between LINC00511 expression and cancers before October 20, 2019, with language limited to English. The terms searched were: neoplasia, tumor, cancer, malignancy, carcinoma, LINC00511, Long Intergenic Non-Protein Coding RNA 511. Details of the search strategy are reported in Inclusion criteria for studies: (1) participants were pathologically confirmed with cancer; (2) the expression of LINC00511 in tissues or plasma was detected; (3) patients with LINC00511 high and low expression group; (4) the relationship between LINC00511 and tumor prognosis  or clinicopathological features was described; (5) relevant data was sufficient to obtain hazard ratio (HR) and 95% confidence interval (CI).Exclusion criteria included reviews, case reports, comments, expert opinions, meta-analysis, and conference summaries. When studies or data duplication were found, only the latest or most complete studies were included.Two reviewers (CC and DXZ) extracted the information about study characteristics, demographic characteristics, the expression of LINC00511 as well as outcomes from remaining studies. If univariate analysis and multivariate analysis are both included in the study, priority is given to extract multivariate analysis result. For studies that did not provide HRs and 95% CIs directly, we estimated them through indirect methods  applied Newcastle Ottawa Scale (NOS), including selection, comparability and outcome, to assessed selected studies critically and independently Stang, . The stuI2 values of 25, 50, and 75% as described previously , a statistical software for data science, was adopted to synthesis and analysis data. The fixed-effect model was chosen to synthesis highly homogeneous data, whereas, the random-effect model and subgroup analysis were conducted for heterogeneous data. Moreover, sensitivity analysis was conducted to assess heterogeneity. We would assign adjectives of low, moderate, and high to GEPIA was performed to obtain the information of LINC00511 across cancer and adjacent normal tissues and relevant prognostic information. Then the expression of LINC00511 among several cancers was plotted and OS and DFS were evaluated. LnCeVars, a comprehensive database synthesizing published studies and abundant datasets , albeit with mild heterogeneity   .I2 = 0.0%, Ph = 0.808) (P = 0.008) (P = 0.754) (P = 0.082)  . Though = 0.008) . The Beg= 0.754) . Moreove= 0.082) .P < 0.001), digestive system cancer , and cervical cancer  (P = 0.001) or indirectly  (P < 0.001) or groups with sample size <100 patients  (P < 0.001) and, to a larger extent, in short follow-up time   . After s< 0.001) . We also< 0.001) . Subsequ< 0.001) . Subgrou< 0.001) .P = 0.27)  (P = 0.008) (P = 0.001) (P < 0.001) (P < 0.001)  . The samctively) . The old= 0.008) . The pat= 0.001) . Patient< 0.001) . Moreove< 0.001) . There w< 0.001) .P < 0.001, We obtained LINC00511 expression across tumor samples and adjacent normal tissues and relevant prognostic information from GEPIA dataset to validate the results. LINC00511 was found over-expressed in most of the cancers, such as liver hepatocellular carcinoma, cervical squamous cell carcinoma and endocervical adenocarcinoma, lung adenocarcinoma, lung squamous cell carcinoma, breast invasive carcinoma, pancreatic adenocarcinoma . Next, wTo further explore the function of LINC00511, genomic variations that disturb LINC00511-associated ceRNA network regulation was curated from LnCeVars dataset. Through gene ontology and functional pathways enrichment analysis, we found that gene ontology was mainly enriched in cell development and death, and apoptosis , and funin vitro; accelerated cell cycle arrest at G0-G1 and apoptosis in vitro; and decreased tumor growth in vivo (Deng et al., LINC00511 shows cancer promoting activity because it was shown that LINC00511 is up-expressed in many cancers and related to several features of cancer, such as proliferation (Sun et al., Moreover, we further strengthened our results with survival information of other types of cancers retrieved from GEPIA. Results from GEPIA were highly consistent with those from meta-analysis. Then, we also conducted a functional analysis of LINC00511-related genes in human tumors to reveal possible mechanisms associated with poor prognosis. We found that gene ontology and LINC00511-associated ceRNA event mainly focuses on the development, death and apoptosis of cells, which is also consistent with the current research reports (Sun et al., Taken all together, the findings presented in this study give a suggestion that LINC00511-mediated diverse pathways might have implications with diverse tumors (Yan et al., Indeed, our study has some limitations. Firstly, we just included 9 studies after strict inclusion and exclusion criteria in the meta section, and some of them are small in sample size. Secondly, all the included studies were conducted in China, which is the reason why we did not conduct ethnic subgroup analysis to investigate any possible effect on the association between LINC00511 and clinical outcomes. Thirdly, some HRs was estimated rather than obtained directly. Finally, all included studies are retrospective, indicating a moderate risk of bias.In summary, elevated LINC00511 is significantly related to age, clinical stage, tumor size, lymph node metastasis, and unfavorable clinical outcome in different types of cancers. Our results showed that LINC00511 may consider as a refreshing biomarker of lymph node metastasis and prognosis in patients with cancer. However, larger sample, multi-ethnic, prospective studies are warranted to confirm the prognostic value of LINC00511. Moreover, LINC00511 acts as ceRNA in the tumourigenesis and development, and this requires further study.http://gepia.cancer-pku.cn/, http://www.bio-bigdata.net/LnCeVar/.Publicly available datasets were analyzed in this study. This data can be found here: X-FW, BL, Y-HZ, and NG conceived, designed, or planned the study. BL and X-FW analyzed the data. CC and D-XZ acquired data. Y-XZ, NS, W-WN, WY, and J-AH helped interpret the results. X-FW and CC provided study materials or patients. D-XZ, BL, and Y-XZ drafted the manuscript. All authors revised and reviewed this work, and all authors gave their final approval of the submitted manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Background and objectives: The association between hypnotic drugs and risk of cancer remains controversial. Therefore, we performed a meta-analysis to investigate this association. Materials and Methods: Pubmed and Embase were searched systematically to identify publications up to April 2020. The Newcastle-Ottawa scale for observational studies was used to assess the quality of studies. All included studies were evaluated by two reviewers independently; any discrepancies were resolved through discussion. Results: Twenty-eight studies including 22 case-control studies and 6 cohort studies with 340,614 hypnotics users and 1,828,057 non-users were included in the final analyses. Hypnotics (benzodiazepines and Z-drugs) use was significantly associated with an increased risk of cancer  in a random-effects meta-analysis of all studies. Subgroup meta-analysis by anxiolytics/sedatives effect (anxiolytics benzodiazepines vs. sedatives group (include sedatives benzodiazepines and Z-drugs)) revealed that a significant association in sedatives group , whereas no significant relationship was observed in anxiolytics benzodiazepines . Moreover, a significant dose–response relationship was observed between the use of hypnotics and the risk of cancer. Conclusions: This meta-analysis revealed association between use of hypnotics drugs and risk of cancer. However, the use of lower dose hypnotics and shorter duration exposed to hypnotics seemed to be not associated with an increased risk of cancer. Moreover, the use of anxiolytics effect benzodiazepines seemed to be lower risk than sedatives benzodiazepines. A high heterogeneity was observed among identified studies, and results were inconsistent in some subgroups. Randomized control trials are needed to confirm the findings in the future. Hypnotics (benzodiazepines and Z-drugs) are medications used to improve sleeping quality and to reduce wakefulness . BenzodiMeanwhile, most common adverse effects associated with hypnotics are residual daytime sedation, drowsiness, cognitive impairment, motor incoordination, dependence ,4,5,6, aCancer is the second leading cause of death globally and leads to economic burden for health systems or costs for patients. Previous studies have assessed the relation between benzodiazepines or Z-drugs use and cancer risk. However, the results are still controversial. In addition, the quantitative meta-analysis has separately examined the benzodiazepines or Z-drugs and cancer risk. Until now, no published quantitative meta-analysis discussed benzodiazepines and Z-drugs long-term use conjunction with cancers. Recently, several observational studies with a larger population indicated conflicting inconsistent results. Combining the results of these studies in a meta-analysis may strengthen their statistical power. In this study, we updated the hypnotic drug correlation between use and the risk of cancer by using a meta-analysis of observational studies both case-control studies and prospective cohort studies.PubMed and EMBASE were searched, by using selected keywords linked with benzodiazepine and non-benzodiazepine hypnotics and the risk of cancer up to April 2020. Keywords were as follows: benzodiazepine or zolpidem or zopiclone or diazepam or alprazolam or clonazepam or temazepam or oxazepam and cancer or tumor or carcinoma or neoplasm and case-control or cohort. Moreover, we reviewed the bibliographies of relevant articles to locate additional studies. The search was limited to human patients and articles in English. The detailed information on the search strategy for eligible studies is given in the flowchart provided by Preferred Reporting Items for Systematic Reviews and Meta-Analyses.The studies we included meet all the following criteria: (1) case-control or cohort study, (2) investigated the associations between the use of benzodiazepines or Z-drugs and the risk of cancer, (3) outcome measures with adjusted odds ratios (OR) or relative risks (RR) and 95% confidence intervals (CI). If data were duplicated or shared in more than one study, the longest-term follow-up studies were included in the analysis. We excluded non-published studies.We used a standardized data extraction form that included study year, study location, study population, participant characteristics, cancer types and crude or adjusted effect sizes and their 95% confidence intervals (CIs).Newcastle-Ottawa Scale (NOS) was used to assess the methodological quality of included case-controls and cohort studies . The quaQ test and I2 statistics were used to assess statistical heterogeneity and inconsistency. Statistical significance was set at p < 0.10 for Cochrane Q tests. Heterogeneity was considered low, moderate, or high, if the 2I values was < 25%, 25–50%, and > 50%, respectively. Results were considered as statistically significant when the p value was less than 0.05. Publication bias was examined by using funnel plots, and Egger’s and Begg’s test was used to analyze the publication bias in our studies. A p-value > 0.05 based on the Egger’s and Begg’s test indicated the absence of publication bias. Statistical analysis was performed according to the Cochrane Handbook for Statistical Review of Interventions (version 5.4) [We conducted meta-analyses on the association of hypnotics use and the risk of cancer by observational studies. The random-effects model (DerSimonian–Laird method) was used to calculate the pooled OR or RR . The CocA total of 658 records were screened, and 113 full-text articles were assessed for eligibility. Twenty-eight articles were selected for qualitative review, including 22 case-control studies and 6 cohort studies . The chaWe performed the methodological quality of studies based on NOS scales. The results of the methodological quality of studies are summarized in p < 0.001, I2 = 79%)  .Subgroup analyses were carried out by research methods design (cohort or case-control study) and hypnotics categories . In the subgroup analyses aligned with study design, hypnotics showed significant positive correlation with risk of cancer, both in case-control and cohort studies subgroup  . In the I2 = 42%), whereas no significant relationship was observed in female with nine studies . However, there was no significant relationship observed in four studies with only elderly subjects . Subgroup meta-analysis by anxiolytics/sedatives effect (anxiolytics benzodiazepines vs. sedatives group (include sedatives benzodiazepines and Z-drugs)) revealed a significant association in sedatives group , whereas no significant relationship was observed in anxiolytics benzodiazepines . Regarding the type of hypnotics, short-acting hypnotics  showed a significantly increased risk. Hypnotics use increased the risk of brain cancer, esophagus cancer, liver cancer, lung cancer, stomach cancer, pancreatic cancer, colon cancer, renal cancer, and prostate cancer. However, no significant association was observed in malignant melanoma, breast cancer, and ovarian cancer. When compared with nonusers of hypnotics, the pooled OR/RR for the risk of cancer was 1.03  in a low dose, 1.30  in a medium dose, and 2.03  in a high dose , and duration of hypnotics use . In recent years, several literatures have reported a tentative link between benzodiazepines and/or Z-drugs exposure with adverse outcomes such as respiratory disease exacerbation, infections, inflammation, dementia, pancreatitis, and cancer . These kAccording to a review by Brambilla et al., the fact that the mechanisms of benzodiazepines- and Z-drugs-induced tumorigenesis remains tentative and unclear . A studyI2 = 84%). It may be associated with the small sample size . Iqbal et al.’s (2014) study uses the Taiwanese National Health Insurance system to gather information about benzodiazepines use and cancer risk. They found that clonazepam, lorazepam, alprazolam, bromazepam, zolpidem, and zopiclone have a high risk of cancer [Currently, there is still a lack of conclusive experimental data, but alarm signals for cancer risk have been raised by researchers for hypnotic drugs based on observational study findings ,27,28,33In addition, the most common risk factors for cancer including aging, smoking, alcohol consumption, family history, and exposure to chemicals or other substances . The stuThis study has some limitations. First, a portion of the included studies in our analysis did not adjust tobacco smoking, alcohol drinking, family cancer history, and medical use factors, which are known as important factors related to cancer. Second, this study only included cohort studies and case-control studies because there are no randomized controlled trials published on this topic. However, cohort studies and case-control studies have a lower level of evidence than randomized controlled trials.The present meta-analysis found that hypnotics use was associated with an increased risk of cancer. However, the use of lower dose hypnotics and shorter duration exposed to hypnotics seemed to be not associated with an increased risk of cancer. Moreover, the use of anxiolytic effect benzodiazepines seemed to be lower risk than sedative benzodiazepines. Further large randomized controlled trials providing a higher level of evidence should be conducted to confirm our findings.


Parental separation is a very common childhood adversity. The association between other adverse childhood experiences and an increased risk of psychosis has been reported. However, the evidence on the risk of psychosis for children of separated parents is limited. In this systematic review, cohort, case–control, and cross-sectional studies, comparing the risk of psychotic disorders for people with and without separated parents, were searched, critically appraised, and summarized.Studies were searched in PubMed, EMBASE, PsycINFO, and the Web of Science, from database inception to September 2019. A meta-analysis, using random-effects models, was undertaken to obtain pooled estimates of the risk of psychosis among participants with separated parents.p < 0.001. The association remained significant when cohort, case–control, and cross-sectional studies were analyzed separately. The five cohort studies included in this review showed and increased risk of psychosis with odds ratio: 1.47 (95% CI: 1.26–1.69), p < 0.001.Twelve studies, with 305,652 participants from 22 countries, were included in the review. A significantly increased risk of psychosis for those with separated parents was observed, with a pooled odds ratio: 1.53 , Parental separation is a common childhood adversity associated with an increased risk of psychosis. Although the risk for an individual child of separated parents is still low, given the high proportion of couple that separate, the increased rates of psychosis may be substantial in the population. Further studies on the risk of psychosis in those with separated parents, and the explanatory factors for this association, are required. The dissolution of marriage is common, affecting 46% of married couples in the USA, 42% in Australia, and 43% in Europe –3. In maParental separation has consequences in children’s life, including exposure to conflict between parents, economic loss, reduced contact with one parent, and increased life stressors such as changing schools, childcare, or homes. Parental separation has become one of the most common adverse childhood experiences, second only to socioeconomic disadvantage ,7. Two sStrong evidence on the association between parental separation and psychosis could identify people at higher risk of having psychotic disorders and inform a proactive and multidisciplinary approach to them. This systematic review and meta-analysis present an up-to-date critical appraisal and summary of the available evidence on the risk of developing psychotic disorders in children of separated parents.The Meta-analysis of Observational Studies in Epidemiology criteria were used to undertake this review . ObservaThe bibliography lists of all papers fitting the inclusion criteria, and relevant reviews identified in the initial search, were checked for further articles. Papers citing all the included studies or relevant reviews were also searched in the Web of Science and considered for inclusion. There were no restrictions on the basis of language, sample size, or duration of follow-up.The following articles were not included: interventional studies; papers that presented only associations with specific domains or outcomes of psychosis; studies with composite outcome  unless separate results for psychosis were presented; those reporting outcome as a continuous variable ; articles presenting only subclinical psychosis or drug induced psychotic disorders; studies comparing participants with different mental health disorders instead of those with and without psychosis; papers in which the exposure was separation between parent and child but parental separation was not one of the reasons for this; those conducted in specific patient subpopulations ; studies reporting results only from univariate analysis; and ecological studies.One researcher (L.A.) conducted all searches of eligible studies. Eligibility was confirmed by two other researchers (M.P.-P. and S.A.). In some cases, the authors of the studies were contacted for any clarifications required including possible publication of two papers from the same sample. Where data from the same participants were presented in more than one paper, only data from the study with larger sample size were included in the analysis. Two researchers extracted the data from all the eligible studies (L.A. and S.A.) using a standardized data-collection form. The risk of bias and overall methodological quality of the studies fitting the inclusion criteria was assessed using the Quality Assessment Tools for Case–Control, Cohort and Cross-Sectional Studies of the National Institute of Health (USA) .p value ≤0.05 was used as a criterion for significance. The heterogeneity between studies was measured using I-squared index (I2), which represents inconsistency of the studies included and estimates the percentage of the total variation due to heterogeneity rather than chance [Pooled estimates of odds ratios (ORs) were obtained using random-effects models . ORs andn chance  All studn chance  and the n chance  were usen chance ,19. To an chance .The electronic search produced 2,210 references, 289 from PUBMED, 960 from EMBASE, 598 from PsychInfo, and 363 from Web of Science. Twelve reviews relevant to the topic, with a total of 1,126 references, and cited by 1,997 papers were also identified ,21–30. TTwelve papers including in total 305,652 participants were considered eligible ,17,31–40The meta-analysis showed that parental separation is significantly associated with psychosis, OR: 1.53 . Even so, all data were checked for accuracy on multiple occasions and all analyses were conducted several times and checked by a senior statistician (S.A.). Finally, some publications may have been miscoded or missed altogether.This study has strengths as well. The comprehensive search and critical assessment of studies conducted in this review allows estimation of the association between parental separation and psychosis obtained on a large number of participants assessed worldwide. The use of a random effect model, based on the assumption that studies were independently conducted and do not necessarily share a common effect size, allowing for more uncertainty of the final summary estimate, was a conservative choice. The overall estimate remained significant despite the increased width of the confidence intervals, providing support to the significance of the findings. The reasonably symmetrical funnel plot suggests provides no evidence of publication bias or small study effects .The risk of psychosis is higher for children exposed to other adversities such as sexual, emotional, physical abuse, or neglect (OR > 2) than for children of separated parents. However, the higher frequency of parental separation can result in a larger number of cases of psychosis associated with it in the population –3. ParenParental separation is associated with an increased risk of psychosis. This should be acknowledged by all professionals working with children of separated parents. Although the risk for an individual child of separated parents is still very low, given the high proportion of couple that separate, the increased rates of psychosis may be substantial in the population. Future studies addressing the possible mechanisms for this association are required.