The effect size was not associated with pre-defined study or patient characteristics, barring the country where the study was conducted. There were no significant differences in effect size between acute myocardial infarction (MI) and unstable angina (UA), and between ST-elevation (STEMI) and non-ST-elevation MI (NSTEMI). However, the effect size was progressively larger in UA (SMD = 1.63), NSTEMI (SMD = 1.91), and STEMI (3.26). Our meta-analysis suggests that IMA might be useful to diagnose ACS. Further studies are warranted to compare the diagnostic performance of IMA vs. established markers, e.g., troponin, and to determine its potential utility in discriminating between UA, NSTEMI, and STEMI (PROSPERO registration number: CRD42021324603).The identification of novel circulating biomarkers of acute coronary syndrome (ACS) may improve diagnosis and management. We conducted a systematic review and meta-analysis of ischaemia-modified albumin (IMA), an emerging biomarker of ischaemia and oxidative stress, in ACS. We searched PubMed, Web of Science, and Scopus from inception to March 2022, and assessed the risk of bias and certainty of evidence with the Joanna Briggs Institute Critical Appraisal Checklist and GRADE, respectively. In 18 studies , IMA concentrations were significantly higher in ACS (standard mean difference, SMD = 2.38, 95% CI 1.88 to 2.88;  Acute coronary syndrome (ACS) and its traditional subtypes, unstable angina (UA), non-ST-elevation myocardial infarction (NSTEMI), and ST-elevation myocardial infarction (STEMI), remain a leading cause of morbidity and mortality worldwide ,2. The eSeveral proteins have been investigated as potential ACS biomarkers in view of their ability to reflect critical pathophysiological processes, e.g., atherosclerotic plaque instability, inflammation, myocardial cell injury, haemodynamic stress, and altered metabolism ,7. One sSerum concentrations of IMA have been detected three hours after the onset of ACS symptoms, with a sensitivity of 70%, a specificity of 80%, and a positive predictive value of 96%, suggesting the potential role of this protein as a biomarker of ACS . TherefoWe searched articles in PubMed, Web of Science, and Scopus, from inception to March 2022, using the following terms: “acute coronary syndrome” or “ACS” or “acute myocardial infarction” or “AMI” or “unstable angina” or “UA” or “non-ST-elevation myocardial infarction” or “NSTEMI” or “ST-elevation myocardial infarction” or “STEMI” and “ischaemia modified albumin” or “IMA”. The abstracts were independently screened by two investigators and, if relevant, the full text was reviewed. Eligibility criteria were: (i) assessment of IMA; (ii) comparison of subjects with or without ACS or its sub-types (case–control design); (iii) use of English language; (iv) availability of the full text. The references of the retrieved articles were also searched to identify additional studies. Any between-reviewer disagreement was resolved by a third investigator. The following information was extracted from each article: age, proportion of males, year of publication, country where the study was conducted, sample size, IMA concentrations, serum troponin concentrations, and ACS subtype. The Joanna Briggs Institute (JBI) Critical Appraisal Checklist for analytical studies was used to assess the risk of bias  . The Grap < 0.05) as different units of measurement  were used. SMD heterogeneity was tested using the Q statistic (significance level at p < 0.10). An I2 value < 30% indicated no or slight heterogeneity, whereas I2 ≥ 30% indicated moderate or substantial heterogeneity [p < 0.05) [Standardised mean differences (SMDs) and 95% confidence intervals (CIs) were used to build forest plots of continuous data, and to evaluate differences in IMA concentrations between participants with or without ACS (significance level at ogeneity . A randoogeneity . Sensitiogeneity . Begg’s  < 0.05) ,19. The  < 0.05) . SubgrouA PRISMA 2020 flow chart is presented in The 18 selected studies included 32 comparator arms in 1654 ACS patients  and 1023 healthy controls or subjects with atypical chest pain without ACS   11,21,2,35,36,37The risk of bias was low in all studies .2 = 97%, p < 0.001). Pooled results showed that IMA concentrations were significantly higher in ACS .The forest plot of IMA concentrations in ACS patients and controls is shown in p < 0.001) but not the heterogeneity .In the sensitivity analysis, the corresponding pooled SMDs were not substantially altered when individual studies were omitted  and Egger’s tests (p = 0.001). The “trim-and-fill” method identified six potential missing studies to be added to the left of the funnel plot to ensure symmetry (p < 0.001).There was a significant publication bias found, according to Begg’s (symmetry . This rep = 0.43), larger in AMI  than UA . Heterogeneity was substantial in both groups (94.4% and 96.8%). Similarly, IMA was able to discriminate between STEMI patients and healthy controls, and between NSTEMI patients and healthy controls (p = 0.42) in effect size between STEMI  and NSTEMI , with substantial study variance in both groups (98.1% and 97.2%). Albeit not significantly, the effect size was progressively larger in UA (SMD = 1.63), NSTEMI (SMD = 1.91), and STEMI (SMD = 3.26). The effect size was also relatively larger in studies conducted in Nepal  and India  when compared to Turkey , Iran  or China . Heterogeneity remained substantial, between 85.0% and 96.9%, in all sub-groups , proportion of males , publication year , sample size , or troponin concentrations . By contrast, a significant association was observed between the effect size and the country where the study was conducted .In univariate meta-regression, there were no significant associations between the effect size and age . After taking into account the low risk of bias in all studies (no rating change), the substantial and unexplained heterogeneity (downgrade one level), the lack of indirectness (no rating change required), the relatively low imprecision , the large effect size , and the presence of publication bias which was addressed with the “trim-and-fill” method (no rating change), the overall level of certainty remained low .Our systematic review and meta-analysis have shown that serum IMA concentrations are significantly higher in patients with ACS, when measured within 24 h of the onset of symptoms, compared to healthy controls or subjects with atypical chest pain without ACS. In subgroup analysis, there were no significant differences in effect size between AMI vs. controls and UA vs. controls, and between STEMI vs. controls and NSTEMI vs. controls. The effect size was progressively, albeit not significantly, larger in UA vs. NSTEMI vs. STEMI, the three traditional subtypes of ACS. Furthermore, the effect size was relatively larger in studies conducted in Nepal and India when compared to those conducted in Turkey, Iran, or China. Barring the country where the study was conducted, in meta-regression the effect size was not significantly associated with a number of study and patient characteristics, including serum troponin concentrations. Therefore, the results of our study suggest that IMA could be a useful biomarker for the early diagnosis of ACS and complement those of a recent meta-analysis investigating the diagnostic accuracy of IMA in ACS. This meta-analysis reported a pooled odds ratio of 3.72, an area under the curve of 0.75, a sensitivity of 0.74, and a specificity of 0.40 .Appropriately designed studies are also warranted to determine the potential utility of IMA in discriminating between UA, NSTEMI, and STEMI, although the routine use of high-sensitivity troponin will likely lead to the incorporation of UA into NSTEMI in the foreseeable future ,4,5. In Several colorimetric and immunochemical methods are available to measure IMA. Some of them, e.g., the albumin copper-binding assay, the enzyme-linked immunosorbent assay, and the surface plasmon resonance immunosensor, are relatively simple and have high sensitivity and specificity . In our p = 0.01) as well as one-year mortality  [p = 0.03) [It is important to emphasise that IMA can also be generated in non-ischaemic conditions that are characterised by various degrees of oxidative stress, e.g., heart failure , neurode= 0.018) . A recen = 0.03) .The strengths of our study include the conduct of subgroup analyses for AMI/UA vs. controls and NSTEMI/STEMI vs. controls, the investigation of possible associations between the effect size and several patient and study characteristics with meta-regression, and the assessment of the certainty of evidence using GRADE. One limitation is that, barring two studies ,24, the Our systematic review and meta-analysis have shown the presence of significant differences in serum IMA concentrations between patients with ACS and healthy controls, or patients with atypical chest pain without ACS. Additional research is warranted to investigate the relationships between IMA generation and the extent of myocardial injury, the effect of revascularization strategies, short- and long-term outcomes, and other specific patient characteristics, including ethnicity. Importantly, these studies should also include patients presenting with typical and atypical chest pain. The results of these studies will determine the potential utility of IMA, singly or in combination with established biomarkers, e.g., high-sensitivity troponin, in the routine diagnosis, risk stratification, and prognosis in patients with ACS.


We conducted this meta-analysis to estimate associations between CDKN2B antisense (CDKN2B-AS) rs2383207 polymorphism and susceptibility to atherosclerosis. A systematic literature research of Google Scholar and PubMed was performed to identify eligible studies. Overall, eight studies were included for meta-analyses. The association was assessed by statistical odds’ ratio (OR) with 95% confidence interval (CI). RevMan software  was used for the meta-analysis. Pooled overall analyses showed that rs2383207 polymorphism was associated with the risk of atherosclerosis in the whole population. Additional analyses by ethnicity revealed that rs2383207 polymorphism was associated with susceptibility to atherosclerosis in Asians and Caucasians. Our results suggest that rs2383207, might serve as genetic biomarkers of atherosclerosis. Further, studies will be required to confirm the observed association. Cardiovascular disease (CVD) is one of the main causes of death in the world. According to data provided by the WHO, 17.9 million people die each year from CVDs, which is about 32% of all deaths in the world. Contributions to the pathogenesis of CVD include genetic and non-genetic factors such as environmental, behavioural and others. Number of premature deaths can be reduced by reducing behavioural risk factors such as smoking, drinking of alcohol, obesity and others.Atherosclerosis underlies cardiovascular diseases and remains a leading cause of morbidity and mortality worldwide. The molecular mechanisms involved in the development of atherosclerosis are not fully understood. Noncoding RNAs (ncRNAs) are transcribed but not translated into proteins, performing their biological functions at the RNA level . AccruinWe have found 122 potentially relevant papers. Among these articles, eight eligible studies were included for pooled analyses . The NOSIn this meta-analysis, 6894 participants were included: 3277 in the group with atherosclerosis and 3617 in the control group. Eight articles were selected: three for rs2383207 in the Caucasian population ,11,12,13Overall allele frequencies were obtained by pooling the alleles of each genotype in focus groups. All investigated polymorphisms contain a major (M) and a minor (m) alleles. So the allele distributions in investigated polymorphisms denominated as MM/Mm/mm.p = 0.007, OR 0.86, 95%CI: 0.78–0.96). No significant associations were observed in overall analyses for other genetic models .To investigate potential associations of CDKN2B-AS rs2383207 polymorphism with susceptibility to atherosclerosis eight studies were included for analyses. Significant association with atherosclerosis was observed for rs2383207 only in heterozygous model . Moreover, rs2383207 polymorphism was associated with susceptibility to atherosclerosis in Caucasians , Further, subgroup analyses by ethnicity revealed that rs2383207 investigated polymorphism were significantly associated with susceptibility to atherosclerosis in Asians  with occurrence of atherosclerosis in Caucasian and Asian populations. Previous studies reported about the association of rs2383207 (A) risk allele with cardiovascular diseases (CAD) , presentp21.3 witThe relationship between CDKN2B-AS and atherosclerosis has been studied intensively. However, no consistent conclusion has been reached. In this meta-analysis, 6894 participants were included; the atherosclerosis and control groups contained 3277 and 3617 individuals, respectively. Statistical significance existed only in heterozygous model comparisons of rs2383207. With the significant outcome in allele comparison, we can come to the conclusion that carrying the mutated allele G on rs2383207 may be associated with atherosclerosis. Subgroup analysis represented potential association between rs2383207 and atherosclerosis for recessive model in the Caucasian subgroup and for homozygous model in the Asian population. However, studies have not included information about influence of environmental factors. Hence, there was no additional analysis of environmental factors such as smoking and drinking alcohol, which may also be associated with atherosclerosis risk.In this mini-review, a potential association between rs2383207 polymorphism in CDKN2B-AS locus and susceptibility to atherosclerosis was analysed. As a result, we have supposed that rs2383207 may be a potential marker for atherosclerosis in Caucasian and Asian populations. This result is quite different from those of previous meta-analyses. Further large scale population based study is needed to consolidate our findings. Additionally, we acknowledge the important role of conducting genome wide association study in identifying the loci that are associated with CVDs. So, research on the influence of CDKN2B-AS with other potential genetics loci, and the environmental factors on CVDs pathogenesis need to be considered with large, representative samples.This meta-analysis conformed with the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guideline . PotentiThe inclusion criteria for publications in this meta-analysis were: (1) studies evaluating the association between the rs2383207 polymorphism and the risk of atherosclerosis using the case–control method; (2) diagnosis of atherosclerosis confirmed by ultrasound evaluation of the carotid arteries; (3) studies with or without Hardy–Weinberg equilibrium (HWE) bias for the control group; (4) DNA copy number was measured using previously described methods based on PCR technology; (5) the sample only included adults over the age of 18; and (6) all data were presented as OR and 95% CI.The main exclusion criteria were: (1) reviews, short communications, conference abstracts, comments; (2) studies without case–control statistics; (3) studies without providing sample data; (4) research in the field of cerebrovascular diseases; and (5) duplicate data.In accordance with the selection criteria, all publications were independently reviewed by two reviewers. The extracted data included basic information: (a) first author name, (b) year of publication, (c) country of origin, (d) ethnicity, (e) genotyping method, (f) sample size, (g) allele and genotype frequencies rs2383207. In addition, we also assessed the methodological quality of the publications included in our meta-analysis using the Newcastle–Ottawa scale (NOS) .p-value < 0.05 was considered statistically significant. I squared statistics were used to assess interstudy heterogeneity. If I squared was higher than 50%, random effects models were used to pool the data. Otherwise, fixed-effect models were chosen. Subgroup analyses by ethnicity of participants and type of disease were also performed. Stabilities of synthetic results were evaluated with sensitivity analyses, and publication bias was evaluated with funnel plots.The data were analysed using Review Manager 5.3 software (RevMan 5.3) to calculate odds ratio (OR) with 95% confidence interval (CI). Heterogeneity was assessed in all studies using the Cochran and I squared tests ,23. The Thus, based on the obtained results, it can be assumed that the CDKN2B-AS rs2383207 polymorphism may be associated with the risk of atherosclerosis among the Asian and Caucasian populations. More studies with larger samples from other ethnic groups are needed to refute or confirm the current findings.


Vitamin E is a group of antioxidative tocopherols and tocotrienols that play a potential role in chemoprevention. Studies investigating the association between vitamin E and prostate cancer risk have been conflicting. We identified observational and interventional studies examining the association between vitamin E intake and prostate cancer risk from PubMed, EMBASE and the Cochrane Library. A random-effects model was used to perform a meta-analysis and estimate relative risks (RRs) and the corresponding 95% confidence intervals (CIs) of prostate cancer risk according to vitamin E intake. Subgroup analyses were conducted by study design, sample size, study population characteristics, geographical region, and dose of vitamin E intake. The association between dietary  and supplemental  vitamin E intake on prostate cancer risk was non-significant. In subgroup analyses, supplemental vitamin E was significantly associated with reduced prostate cancer risk in studies in Europe . Overall, this meta-analysis demonstrates little evidence for a beneficial effect of vitamin E intake on prostate cancer risk but suggests that there may be some conditions in which supplements could confer a protective effect on prostate cancer risk. Prostate cancer is characterised by uncontrolled cell growth within the prostate, which is a small muscular gland located below the bladder in men. In more than half of the countries in the world, prostate cancer is the most frequently diagnosed cancer among men . GloballVitamin E is a group of antioxidant fat-soluble micronutrients  that is found in high amounts (>10 mg per 100 mL or 100 mg) in nuts, seeds and vegetable oil . It has Previously, a meta-analysis of RCTs using vitamin E-containing interventions estimated a significant protective effect for vitamin E on prostate cancer risk . HoweverThe protocol was developed in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA-P) statement .A search of the electronic databases PUBMED, EMBASE, and the Cochrane Library for references published before 30 November 2021 was undertaken. We used a combination of standard controlled vocabulary (MeSH and Emtree) and keywords for (1) vitamin E and (2) prostate cancer to identify relevant references that studied the association between vitamin E and prostate cancer risk. The search strings used for all three databases are presented in We included RCTs and observational studies  that studied the association between vitamin intake  and prostate cancer risk in our meta-analysis. RCTs evaluating the use of a multi-vitamin supplement were eligible if vitamin E was described as a component. Studies had to have enrolled adult male participants over 40 years of age and presented risk estimates of prostate cancer  as well as the respective 95% CI according to categories of dietary and/or supplemental vitamin E intake.If multiple studies used data from the same population, the study with the longest follow-up time was included in this meta-analysis. Included studies had to be written in English and published in peer-reviewed journals.Two reviewers, LWQ and YJY, independently screened all references downloaded from the three databases. The following data items were extracted from the included studies: publication information , study population , exposure or intervention of vitamin E intake , prostate cancer risk estimates of highest versus lowest category of vitamin E intake , as well as any covariates that were adjusted for in the models. When results from multiple statistical models were presented, risk estimates and the respective 95% CI were extracted from the model that adjusted for more covariates. Data extraction was completed through an electronic form developed in Microsoft Excel. Any inconsistencies between the two reviewers were resolved through discussion and consensus.Two reviewers, LWQ and YJY, evaluated the risk of bias of individual studies independently. RCTs were assessed using the Revised Cochrane Risk-of-Bias (RoB 2) tool, which consists of five domains: (1) bias arising from the randomization process, (2) bias due to deviations from intended interventions, (3) bias due to missing outcome data, (4) bias in the measurement of the outcomes, and (5) bias in the selection of the reported result . The risObservational studies were assessed using the Cochrane tool Risk of Bias In Non-randomised Studies of Interventions (ROBINS-I) . This to2 values. Synthesis of the overall estimate was done with careful consideration of the heterogeneity, making sure that studies were only combined under appropriate heterogeneity levels (<70%). Subgroup analyses according to study design, sample size, study population characteristics, geographical region and dose of vitamin E intake were conducted to determine if pooled risk estimates differed across these characteristics. To assess the robustness of our findings, we performed sensitivity analyses by excluding one study at a time to examine the magnitude of influence each study had on the pooled risk estimates. We also carried out further sensitivity analyses by excluding studies with a high or serious risk of bias, as determined by the RoB2 and ROBINS-I tools, as well as limiting studies to those that had adjusted for important confounders . We also attempted to estimate the trend of relative risk estimates across increasing categories of vitamin E intake through a dose-response meta-analysis using the ‘dosres’ package in R [Using data extracted from the included studies, pooled risk estimates and respective 95% CIs for prostate cancer risk according to vitamin E intake (highest category versus lowest category) were computed. We chose to implement the random effects model in this meta-analysis as the studies were clinically heterogeneous , and were likely to estimate different underlying true effects. The restricted maximum likelihood method was used to estimate the heterogeneity variance. We examined and quantified the statistical heterogeneity of the study results using a Cochrane Q test and Iage in R ,24,25.p values were two-sided, with a value of <0.05 being considered as statistically significant.We examined small study effects, including potential publication bias, by carrying out a funnel plot analysis and assessing asymmetry using an Egger’s regression test. The trim-and-fill method was applied to calculate an adjusted effect size. All analyses were done using the ‘metafor’ package in R  . All p vDetails of the study selection process are presented in a PRISMA flowchart . A totalWe identified 19 studies that investigated associations between dietary intake of vitamin E and prostate cancer risk. This included 11 case-control studies, six prospective cohort studies, one case-cohort study and one nested case-control study. In total, the analysis comprised 516,753 participants, in which 27,141 cases of prostate cancer were identified.Meanwhile, we identified a total of 18 studies that evaluated the supplemental intake of vitamin E and prostate cancer risk, comprising six RCTs, nine prospective cohort studies, two case-control studies and one case-cohort study. Here, the analysis consisted of 686,348 non-overlapping participants, among which 31,274 cases of prostate cancer had been identified.There were five studies evaluating the association between total  intake of vitamin E and prostate cancer risk. This comprised one case-control study, three prospective cohort studies and one case-cohort study. In total, the analysis included 94,421 non-overlapping participants, in which 5122 cases of prostate cancer had been identified.2 = 7.47%; 2 = 0.00%; 2 < 25%) or moderate (25% ≤ I2 < 50%).The pooled relative risk estimates of prostate cancer risk were 0.97  and supplemental intake  and prostate cancer risk. Applying the trim-and-fill method did not drastically alter the risk estimates . We did not carry out an assessment of small-study effects for studies on total vitamin E intake as the number of studies with this variable were limited.To assess the presence of small-study effects, we carried out funnel-plot analyses of the included studies on dietary A and supThe overall risk of bias for RCTs and observational studies are presented in 2 = 0.04%). Further sensitivity analyses by excluding observational studies at ‘serious’ risk of bias or RCTs with ‘some concerns’ of bias did not alter our findings for dietary vitamin E  or supplemental vitamin E . In addition, restricting studies to those that had adjusted for energy intake  or family history of prostate cancer ; supplemental vitamin E: RR = 1.00; 95% CI = 0.95–1.04; I2 = 0.00%) did not alter our findings. There was limited evidence of a dose-response relationship for dietary or supplemental vitamin E intake and prostate cancer risk. Sensitivity analysis for studies on total vitamin E intake were not conducted, as the number of studies on this variable were limited.During the leave-one-out sensitivity analysis, pooled relative risk estimates remained largely similar to original estimates for dietary/supplemental intake of vitamin E and prostate cancer risk. However, when excluding the prospective cohort study by Stram et al., the inverse association between dietary vitamin E intake and prostate cancer risk attained borderline significance .Results of the subgroup analyses are summarised in 2 = 33.56%) and North American populations  were significantly different . However, study design, study type, sample size, dose of supplements used, as well as the presence of underlying conditions in participants of RCTs, did not appear to influence estimates of prostate cancer risk with supplemental vitamin E intake.For supplemental vitamin E intake, the pooled risk estimates in studies of European populations  were suggestive of a slight reduction in prostate cancer risk for high dietary intake of vitamin E, there was no statistical evidence for a benefit. This implies that a high dietary intake of vitamin E may not account for the inverse associations between serum/plasma vitamin E  and prostate cancer risk that have been reported in the literature [Although the risk estimates  in evaluating this exposure variable.2 values for the overall analysis were not substantial (<40%), indicating that the studies included could be reasonably combined. Furthermore, our subgroup analysis by study design suggested that there were no differences in effect detected between studies of various designs.One strength of our meta-analysis is that a low level of heterogeneity was observed within the overall analysis. Results obtained by combining studies of different designs may lead to bias resulting from heterogeneity. the ISecondly, our study took into account both the dietary and supplemental intake of vitamin E, as the effect of these intakes are likely to be interrelated. Past meta-analyses have focused on the use of vitamin E supplements alone and prostate cancer risk, or combined a mixture of studies on either dietary and supplemental intake, without considering that the exposure variables were distinct but also additive ,16,62.However, there were also several limitations with our meta-analysis. Firstly, the studies used various assessment methods to measure intake, as well as different cut-off values to categorize intake. These variations introduce heterogeneity and make the pooled risk estimates difficult to interpret. Although a dose-response meta-analysis could have addressed this by enabling one to extract multiple risk estimates across varying doses from the same study, we were limited by the number of studies providing sufficient information  for such analysis. We also acknowledge that the lack of association observed in the meta-analysis could be due to a lack of statistical power or a lack of precision due to the inherent information biases of measuring dietary intake through methods such as food-frequency questionnaires and food diaries. Here, non-differential misclassification may bias the results towards the null.Secondly, we were unable to factor in the bioavailability of vitamin E in our meta-analysis. The absorption rate of vitamin E, which can range from 20% to 80%, is heavily influenced by other factors such as the food matrix present . We wereFurthermore, most studies in the meta-analysis do not differentiate between the subtypes of vitamin E. This may be limited by the challenges faced in separation and quantification of tocopherols and tocotrienols in various food types . As a reAdditionally, the studies included in our meta-analysis tended to overlook fortified sources and cooking oils, which could be major sources of vitamin E when computing intake. This could have led to an underestimate of vitamin E intake and/or non-differential misclassification, which may have biased the results towards the null.Lastly, as the majority of the studies had been conducted in Western populations in Europe or North America, the findings may face limited generalisability to populations from other regions, such as Africa or Asia.Overall, our meta-analysis implies that neither vitamin E supplements, nor a high dietary intake of vitamin E, are likely to be significantly useful in a chemopreventive capacity for prostate cancer. However, future studies may benefit from investigating the intake of specific vitamin E forms and prostate cancer risk in various geographical regions.


Objective: It is suggested that estrogen receptors (ERs) might be associated with the disproportionate vulnerability of women to depressive episodes. Several variants in ER-alpha (ERα) and ER-beta (ERβ) have been linked to depression, but the results were not consistent. Hence, we conducted a meta-analysis to evaluate the association between ERα/ERβ and depression in a cohort of women.Methods: A comprehensive literature search was performed in public databases. The genetic association between polymorphisms in Erα/ERβ and depression risk in a cohort of women was evaluated by odds ratios (ORs) and 95% confidence intervals (CIs). Cochran’s Q test and the I2 index were used to evaluate heterogeneity.Results: In total, 10 studies and 4 SNPs  were included in our meta-analysis. rs2234693 genotype was significantly associated with the risk of depression in women by dominant model , recessive model , additive model  and allelic model . For rs9340799, the frequencies of risk genotypes according to the dominant  and allelic  models were significantly lower in women with depression than in controls within the Asian subgroup. For rs1256049, risk genotypes were significantly more frequent in depressed subjects than in controls under the dominant model  and the allelic model  after sensitivity analysis by omitting one study which induce the heterogeneity.Conclusions: The current meta-analysis is the first and most comprehensive investigation of the association between ERs and depression in women, and the findings support the concept that ERs participate in the etiology of sex heterogeneity in depression. Depression is characterized by a persistent depressed mood and/or loss of pleasure in activities . There iThere is a great deal of evidence that the sex heterogeneity of depression is largely due to estrogens and estrogen receptors (ERs) . WithdraThe cellular effects of estrogens are mainly due to the activation of two estrogen receptors, known as ER alpha (ERα) and ERβ . These tTo date, several variants in these two receptors have been associated with depression, especially in women; these variants include rs2234693, rs9340799, rs4986938 and TA repeat in ERα as well as rs1256049 and rs2077647, rs1271572 in Erβ . Howeverhttp://www.cochranehandbook.org) and the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement  from inception to 2 March 2022. No limitations were imposed on publication year or language. The search terms used to identify studies were a combination of terms identifying the genes  and terms identifying the phenotypes of interest . Studies that met the following criteria were included in the analysis: 1) the study was performed on adult women (≥18 years old) with a diagnosis of depression; 2) the study reported data from at least one independent sample; and 3) the study provided sufficient data to calculate the odds ratios (ORs) and their 95% confidence intervals (CIs). Studies were excluded if 1) they were performed on youth under 18 years old or on men, or 2) there was no control group. The reference lists of all primary studies were also reviewed. For multiple publications covering the same sample, the one with the most complete and recent report was included.Two reviewers independently extracted the data and study characteristics, including the following: first author name and publication year, study design, participant information , diagnostic criteria, assessment, genotyping methods, genotype and allelic distribution.A quality assessment was performed according to the guidelines laid out in the HuGE Review Handbook . We applI2 index based on a chi-square distribution were applied to estimate heterogeneity. I2 ranges from 0 to 100%. Low, moderate, high and extreme heterogeneity correspond to0-25%, 25–50%, 50–75% and75-100%, respectively . Sensitivity analysis for the overall effect was performed by omitting one study at a time. Potential publication bias was evaluated by Begg’s funnel plots, Egger’s  and R Studio version 1.2.1 .HWE was tested using Pearson’s chi-squared test. Four genetic models were used in the meta-analysis: the dominant model, the recessive model, the additive model and the allelic model. ORs with 95% CI were used to assess the different distributions of ER gene polymorphisms in the depression and control groups within this cohort of women. Cochran’s Q test and ectively . If the  Egger’s  and Pete Egger’s  regressiIn total, 2566 articles were obtained from the initial search. The flow diagram of the process of selection is illustrated in p = 0.0031 under a fixed-effects model), the recessive model , the additive model  and the allelic model   and the additive model . The sensitivity analysis was performed, and the data showed that no individual study qualitatively altered the pooled ORs in the four models  . The hetr models . This inr models .p = 0.0405) with low heterogeneity   and allelic model  under the random-effects model, but still with a large heterogeneity  and allelic model  under the fixed-effect model, but the same was not true of the recessive model or the additive model  under a fixed-effects model  . There w = 0.55) . Due to ogeneity . The senve model . There wve model .2 = 59%, p = 0.05) and allelic model . The most effective reduction in heterogeneity  emerged when Kang’s study was omitted  and the allelic model   . There wThe meta-analysis failed to find an association between depression susceptibility and the rs4986938 polymorphism  in womenAccording to recent studies, ERα and ERβ polymorphisms are believed to be closely involved in sex-specific clinical symptoms and outcomes in women with depression . This suThe rs2234693 polymorphism is located in the first intron of ERα and has been reported to impact the response of the ERα gene to estrogen by altering transcription factor binding . Many re2 = 0%). The results were still significant in the pooled sample from the three remaining studies. In the Caucasian subgroup, we failed to find any significant difference in the pooled data from two studies. Collectively, the G allele of the rs9340799 polymorphism is a potential risk allele for depression in Asian women.The ERα rs9340799 polymorphism has been widely studied, but conflicting results have been reported, and high heterogeneity was found in our meta-analysis. The variations and heterogeneity are particularly relevant in genetic heterogeneity in ethnicity. Therefore, subgroup analyses based on ethnicity were conducted in the present meta-analysis. In the Asian subgroup specifically, the G allele showed a tendency to be a risk factor for depression in women, and significant ORs were reported in the dominant model and the allelic model of rs9340799, albeit with high heterogeneity. The high heterogeneity might derive from the unstable allelic frequency, which was induced by the limitation of the sample size in individual studies. For instance, in the patient group of a Korean sample with 43 female depression patients and 63 controls, Kim et al.  noted a rs1256049 is a silent mutation in exon 5 of ERβ gene. Zhang et al.  reportedThe limitations in this meta-analysis should be noted. For example, because of the limited effective number of samples, our meta-analysis was unable to test the relationship of ERα and ERβ polymorphisms with depression susceptibility in the Caucasian subgroup. Additionally, our meta-analysis pooled samples of women with depression in the gestational, menopausal, perinatal and other periods. Further subgroup analysis should be carried out based on these different periods marked by distinct hormonal fluctuations, such analysis may help to clarify the role of hormones in depression and contribute to effective treatment options.In conclusion, our meta-analysis yielded evidence that the ERα polymorphism rs2234693 is associated with susceptibility to depression in women. Additionally, the ERα rs9340799 polymorphism is a potential risk factor for depression under the dominant and allelic models in Asian women. Further studies with larger sample sizes are still needed to support this conclusion.


The paper outlines several approaches for dealing with meta-analyses of count outcome data. These counts are the accumulation of occurred events, and these events might be rare, so a special feature of the meta-analysis is dealing with low counts including zero-count studies. Emphasis is put on approaches which are state of the art for count data modelling including mixed log-linear (Poisson) and mixed logistic  regression as well as nonparametric mixture models for count data of Poisson and binomial type. A simulation study investigates the performance and capability of discrete mixture models in estimating effect heterogeneity. The approaches are exemplified on a meta-analytic case study investigating the acceptance of bibliotherapy.The online version contains supplementary material available at 10.1007/s11336-021-09835-5. This setting can be described by a random count variable i indicates the study i for k denotes the number of available studies. Also, i and group j, whereas i and group j. The latter is considered as non-random and is also called the number at risk.Meta-analyses are used to analyse and integrate the results of several studies investigating the same research question, providing a less costly and more powerful alternative to a large new single study. For a general introduction into meta-analysis, refer to Borenstein et al.  or SchulA conventional two-stage meta-analysis proceeds as follows. In the first stage, an estimate of an effect size such as the relative risk The meta-analytic one-stage approach is closely connected to multilevel analysis and modelling as studies introduce a natural hierarchical level in the data. As pointed out in Hox et al. , meta-anRare events meta-analysis experiences serious drawbacks if conducted following a conventional pathway. Effect measures such as risk or odds ratio might be undefined, as would be the associated approximate variance estimates, unless continuity corrections are invoked with unclear bias potential. In addition, the within-study normality assumption for the effect measure might be seriously in doubt. Here, it is suggested to use count model approaches such as generalised linear and generalised mixed models, as they have been developed and well-investigated for counts and found to perform considerably well.More importantly, finite mixture models are suggested to replace the parametric  random effects distribution. We see this as an important step towards creating a new generation of two-level nonparametric meta-analytic approaches. Here we propose to allow mixing on baseline and, potentially, on the effect parameter itself. In addition, we demonstrate in simulation work that these methods can be used successfully in identifying the underlying risk structure.The paper is organised as follows. Section In summary, we outline the major novel aspects of the paper in the following:We use meta-analytic data on the acceptability of bibliotherapy compared with control groups in the treatment of children and adolescents with depression and/or anxiety from eight studies to illustrate the application of the risk ratio and odds ratio. These data were provided by Yuan et al. . Here, bp-value of 0.14 for the risk ratio and 0.12 for the odds ratio), the estimated overall risk ratio and overall odds ratio are given by 1.86 and 2.08, respectively. These are obtained with the Mantel–Haenszel estimator which allows zero-containing studies, and of which details are given in the following section. It can therefore be hypothesised that all-cause discontinuation is observed more often in bibliotherapy than in control conditions, indicating lower acceptability of bibliotherapy in children and adolescents with depression and/or anxiety. This question will now be further investigated in the following sections.Meta-analytic data of bibliotherapy and control conditions for acceptability used in this example are given in Table In the following, we look at the relative risk and odds ratio, and how they generalise to log-linear and logistic regression. Details can be found in Jewell , for exai, the relative risk is modelled using the log-linear modeli, as can be seen by substituting in empirical risk ratio i, assuming that Conditional upon study j=1 into . If we ri, the odds ratio is modelled using the logistic modeli, as can be seen by substituting in A second measure is the odds ratio. Here, conditional upon study j=1 into .empirical odds ratio i, assuming that i and Again, if we replace the expected values by their observed counterparts, we find the effect homogeneity. In other words, we consider that i. We would still like to keep a specific intercept value baseline heterogeneity as it refers to the risk or odds in the control group, which represents the baseline population. Two approaches are possible. In the first approach, a parameter estimate i. This is sometimes called a contrast-based approach. In the second approach arm-based approach.Here, we consider a specific case of models  and 2),,2), nameWe emphasise here the difference between a contrast-based approach and an arm-based approach–which refers largely to the way the intercept is modelled. In an approach, where the study factor is ignored, there is high potential for confounding. We illustrate this issue in the discussion with a synthetic example. As there is agreement that the study factor should be adjusted for, the question arises how this can be best accomplished. One opinion is that one should avoid situations where interventions are compared across studies, which can be achieved by entering study as a main effect. The concern here is not whether the main effect parameters of the study factor (baseline parameters) are estimated validly (a Neyman–Scott problem may occur referring to the issue of consistently estimating infinitely many nuisance parameters), the question is whether we can achieve an unconfounded estimate of the effect of interest. Another opinion is that an unconfounded estimate of the effect can be achieved by treating the main effect of study as a random effect . We continue here with both approaches and will see that they both lead to identical effect estimates.i and We also mention here the Mantel–Haenszel approach  and Bayesian information criterion (BIC) for each model considered here. The overall message here is that all approaches perform similarly. Estimating log-linear and logistic models with fixed and random intercepts, i.e. without and with baseline heterogeneity, we find that the risk of all-cause discontinuation is significantly increased by a factor of 1.84 for bibliotherapy and that the odds for all-cause discontinuation are significantly increased by a factor of 2.09 in bibliotherapy. In addition, confidence intervals are fairly similar.Table To model effect heterogeneity of the effect measure In Table For comparison, we included in Table In the following, we lay out how a nonparametric random effects approach can be developed using nonparametric mixture models. Key theoretical results can be found in Lindsay , whereasith study is given by baseline heterogeneity—the variability in the intercept QS should be chosen. This is known as the number of components problem. A typical solution is to start with S, the log-likelihood  at the University of Münster. Computations were parallelised using the doParallel package  and Bayesian Information Criterion (BIC) were used to determine the preferred model, thereby taking into account both model fit and model complexity. The AIC and BIC are widely used criteria for model selection  the number of studies, (ii) the size of samples within studies, (iii) the component weights Just like for the first simulation study, the results of the second simulation study were evaluated in terms of model selection performance using the AIC and BIC, as well as performance with regard to the estimation of In this paper, we presented alternatives to conventional two-stage approaches for meta-analysis from the family of generalised linear mixed models and nonparametric mixture models. These alternatives overcome the shortcomings of conventional inverse variance-weighted two-stage models, where studies with zero-counts cannot be included when adding a smoothing constant. In addition, they allow separate modelling of baseline heterogeneity and effect heterogeneity, while the Mantel–Haenszel approach rests on the assumption of homogeneity. In contrast with the log-linear and logistic mixed models presented, nonparametric mixture models do not require specification of the random effects distribution. In particular, these models avoid the assumption of a normal random effects distribution, as this assumption cannot be easily investigated. This is also correct for alternative random effect distributions such as the Gamma or the Beta distribution, as these are mainly chosen for mathematical convenience as they allow closed form solutions for the marginal integrals . The nonparametric approach circumvents this issue entirely.We return to the issue that consideration of the study level variation can be crucial, and simply pooling the data across studies would have considerable confounding potential. Let us consider the following synthetic example. We generate two types of studies. In study type A, we have a baseline risk of 0.5 with 100 persons at risk in the control group. Ten persons are at risk in the treatment group where the risk is also 0.5, so that the risk ratio is 1. In study type B, we have a baseline risk of 0.1 with 10 persons at risk in the control group. One hundred persons are at risk in the treatment group for which the risk is also 0.1. In both studies, the risk ratio is 1; in other words, there is no effect present. We generate 20 studies of type A and 20 of type B, to keep the scenario realistic. Common sense would tell us, independent of whether you favour the arm-based or contrast-based approach, that any decent analysis would come to the conclusion of no effect. The pooled analysis provides a risk ratio of 0.28 with 95% confidence interval , so a clear and significant effect. Including the baseline parameter as a fixed main effect in the Poisson model yields a risk ratio (CI) of 1.01  and the Mantel–Haenszel estimate is 1.01 , identical to the former up to two decimal places. Including the baseline parameter as a random effect yields a risk ratio of 0.83 , also avoiding the strong confounding effect of the unadjusted effect. Of course, the case study was constructed to make this point, by choosing strong baseline risk variation and highly unbalanced intervention and control groups in the studies, which are typical conditions for the occurrence of confounding. Another possibility would be to eliminate the baseline nuisance parameter prior to any further modelling. This approach is laid out in detail in Böhning et al.  for the A challenging situation for meta-analysis is when the outcome of interest has very low probability. Some tables may have empty cells for one or both treatments.Bonett and Price (We consider the risk ratio (or odds ratio) in our setting. Although the risk difference has considerable benefits and is indeed statistically easier to treat, it is not without reason that the statistical  community often favours the risk (odds) ratio. One of the major reasons is that the latter is invariant towards study duration , whereas this is not the case for the risk difference. In other words, the risk (odds) ratio is a relative effect measure, whereas the risk difference is an absolute one. Alternative meta-analytic approaches for the risk and odds ratio were laid out by Bonett and Price , 2015. Tnd Price , Bradburnd Price , Kuss  comparing a psychological intervention with control conditions with outcomes such as complete recovery, remission, response, dropout rates, acceptability or adverse events. For such meta-analyses, it can be beneficial to separately model baseline heterogeneity and effect heterogeneity: RCTs differ in their types of control conditions , which makes it reasonable to expect heterogeneity in the baseline population. Also, although inclusion criteria for the treatment condition are often more narrowly defined, there are still various methodological aspects which can differ across studies, such as treatment duration and intensity, treatment material, and whether patients with comorbidities were included or not, which potentially introduces effect heterogeneity. Both generalised linear mixed models and nonparametric mixture models allow for such heterogeneity to be accounted for.lme4), software packages for nonparametric mixture models (such as the flexmix package) do not enjoy a similar degree of popularity. Also, there is a lack of simulation studies comparing nonparametric mixture models to generalised linear mixed models. The latter were found to perform well in various simulation studies and for different meta-analytic settings Below is the link to the electronic supplementary material.


Epidemiologic studies show a strong association between chronic inflammation and type 2 diabetes (T2D). Diet may also affect the risk of T2D by modulating inflammation. This meta-analysis aimed to assess the relation of dietary inflammatory index (DII) and risk of T2D.  PubMed and Scopus were systematically searched from their inception to September 2020 to identify relevant studies. Relative risks, hazard ratios, or odds ratios (OR), with their corresponding 95% confidence intervals (95% CI), were calculated and pooled using a random-effects model. I2 = 96.5%, P < 0.001); however, higher DII was identified as being significantly related to increased risk of T2D in high quality studies . In the stratified analysis by the dietary assessment tool, background disease, and sex of participants, DII showed no significant association with T2D.  A total of 48 different studies, with a total sample size of 1,687,424 participants, were eligible to be included in this meta-analysis. In the overall analysis, no significant association was observed between DII and risk of T2D , with significant evidence for heterogeneity ( Higher DII might be associated with an increased risk of T2D. Additional well-designed studies are required to confirm this finding. No limitation was placed on date, and all studies from Medline inception to January 2021 were imported to an EndNote library for screening. Search was conducted by SA.This review included eligible observation studies  examining potential inflammatory of diet (exposure) as a function of T2D (outcome) and reported relative risks (RR), hazard ratios (HR), or odds ratios (OR) or provided sufficient data to calculate them; when data were not reported in the studies, we contacted the corresponding author to obtain data. Criteria for exclusion included the following: clinical trials, editorials, books, dissertation, and other grey literature. Data were extracted independently by two reviewers, and disagreements were resolved by discussion.All the records were imported to an EndNote library for enhancing the screening process. AM and HM screened titles and abstracts for potential eligibility and removed records based on predefined criteria. In studies with T2D as the secondary outcome, it was difficult to determine eligibility based on title/abstract, so full text of 80% of studies were retrieved for further inspection. Retrieved papers were read in full and all references followed up.An evidence table with a priori determined headings was used for data extraction of eligible studies as follows: the first author's name, geographical setting and population, year of publication, sex and mean age of participants, sample size, DII scoring system(s) and assessment tool(s), T2D assessment tool, DII comparison cut-points, documented effect sizes, and covariates adjusted, if available. Data were extracted independently by two reviewers, and in case of disagreement, a third reviewer was involved.∗”: 7–9 ; 4–6 ; 1–3 .Two reviewers (SA and HM) separately filled out Newcastle–Ottawa Scale (NOS), one of the best tools for quality appraisal of observational studies, for included studies. Cross-sectional and case-control studies were assigned a maximum of 10 asterisks for the three domains of selection (0–4), comparability (0–2), and assessment of exposure (0–3). Each cohort study was also given a maximum of 10 asterisks for parameters in the three domains of selection (0–5), comparability (0–2), and assessment of outcome (0–3). Final quality scores were determined based on the number of criteria received “I2 values). I2 > 50% together with P < 0.10 was used to determine statistically significant heterogeneity. Subgroup analyses using random-effects models investigated potential sources of variation that were due to reasons other than chance. Potential sources examined were quality of studies, dietary assessment tool, and sex of participants. The possibility of publication bias was assessed by conducting Egger's test  and constructing funnel plots for visual inspection. Metaregression analysis was conducted to assess if pooled estimates were affected by age of participants. Sensitivity analysis was conducted to assess the robustness of the synthesized results. All analyses were done using Stata, version 13 . Pooled estimates' significance was set at P < 0.05.Studies reporting the association of DII with T2D using odds ratios, hazard ratios, or relative risks were considered as eligible for qualitative synthesis. However, for quantitative synthesis, log-transformed risk ratios as well as corresponding standard errors (SEs) were obtained from ORs, RRs, or HRs extracted from models that were adjusted for the highest number of covariates. If the study did not report these three risk ratios directly, we used the raw data  provided by the study to estimate the unadjusted ORs. Due to great variety in the included studies, particularly in terms of DII scoring system, a random-effects model was used for pooling risk estimates on a forest plot regardless of the presence or absence of heterogeneity (n = 55) and irrelevant publications (n = 310) based on title and abstract, 56 articles underwent full-text screening. Among them, 8 further studies were excluded since they were reviews, had irrelevant exposure or outcome, or reported results according to the linear regression analysis (beta coefficient), and thus their risk estimates were not calculable [The preliminary search through databases yielded a total of 421 publications. After excluding duplicate ; however, higher DII was identified as being significantly related to a 58% increased risk of T2D in high quality studies   . No signto 2.17) .t = 0.62, P = 0.54)   . Further = 0.16) .Sensitivity analysis by omitting one study at a time did not significantly change the results, showing the reliability of the findings .Studies exploring the relation of DII to the risk of T2M have yielded inconclusive findings. This meta-analysis aimed to resolve these inconsistencies by pooling the available observational studies. The pooled effect size of high quality studies showed that higher DII is significantly related to the increased risk of T2D, although this finding was not supported in the stratified analyses by dietary assessment tool and sex of participants.Long time exposure to low-grade inflammation could result in an elevation in the risk of obesity and its-related metabolic complications such as T2D . The relγ, IL-1, IL-6, IL-8, CRP, and TNF-α), which can lead to insulin resistance [The precise mechanism by which diet-related inflammation might elevate the odds of T2D is not well known although it is well established that T2D is fostered in a proinflammatory setting . A proinsistance . It has sistance . In thissistance –84 and Tsistance , 86. Howsistance , 88, possistance . In addisistance , 52, butsistance , have resistance ; on the sistance  in proinsistance  and imprsistance , justifyThe present study is the first meta-analysis to comprehensively investigate the relation of DII to T2D. The results have significant clinical implications as targeting the DII components could be a useful strategy to decrease the risk of T2D. As a strength point, there was no significant evidence for publication bias. However, some limitations of our meta-analysis should be considered. First, T2D was a secondary outcome in the majority of the included studies, and we calculated crude effect size for such studies, which may increase their susceptibility to bias. Nevertheless, to resolve this issue, we performed subgroup analysis based on the study quality to obtain a more reliable conclusion. Second, a significant heterogeneity was detected across the publications; despite that, we applied random-effects analysis to consider the observed heterogeneity. This heterogeneity can be derived from differences in genetic background, background disease, level of adjustment for covariates, sample size, and quality of studies. Subgroup analysis found that dietary questionnaire used to calculate DII was a potential source of heterogeneity, showing that differences in dietary assessment tool among previous publications may partly justify the detected heterogeneity. Third, the analyzed publications were observational in design, which are subject to selection bias, and causality could not be inferred. Finally, number and type of food components applied for the computation of the DII varied across studies, and standardizing the intake of each food components according to world mean and standard deviation (SD) complicates the comparability of DII across different populations, which may be a source of inconsistent findings in the prior studies.In conclusion, this meta-analysis indicated that adherence to a proinflammatory diet may increase the risk of T2D. Thus, the recommendation of a healthy dietary pattern may decrease diet-induced inflammation and subsequently lower the risk of T2D. Additional studies are required to identify whether a diet that particularly targets the DII components could be useful clinically to decrease the development of T2D.


In addition, subgroup analyses were performed to reduce the heterogeneity of the initial meta-analyses. The results provided strong evidence that inhaling radon in the indoor environments is closely associated with the development of lung cancer and childhood leukemia in patients living in Europe and areas with high radon levels (≥100 Bq/m3).This study investigated the impact of residential radon exposure on human cancers  through a systematic review and meta-analysis of case–control studies. A total of 9724 articles obtained from electronic databases were assessed; however, only 55 case–control studies were eligible after manually screening and eliminating unnecessary studies. The causal associations were addressed by determining the meta-analysis’s estimated size effects  of the meta-analysis. Residential radon was revealed to significantly increase the incidence of lung cancer and childhood leukemia with pooled ORs of 1.38  (I However, the heterogeneity was primarily high owing to the significant variation in the individual study regions, sample size, and other effect factors .In the sensitivity analysis, removing one study at a time did not significantly affect the pooled OR; it ranged from 1.30  (I2 = 0% and p = 0.51). In contrast, the pooled RR suggested no association between radon exposure and cancer . The heterogeneities did not exist in the comparison based on OR values, while the comparison based on RR values exhibited a moderate degree of heterogeneity (I2 = 0% and 52%).The pooled estimate of 13 case–control studies for the increase in childhood leukemia induced by residential radon exposure was divided into subcomparisons based on the estimated size effects, namely, OR and RR . The res3, 100–150 Bq/m3, and ≥200 Bq/m3. Finally, the included studies were categorized regarding smoking status of patients, including nonsmokers and smokers.In this study, a subgroup analysis was designed to assess the impact of residential radon inhalation on human health based on estimated OR values corresponding to four subcriteria . First, 2 > 50%). In the analysis of the study population, a statistically significant increase in lung cancer due to residential radon exposure was observed in the case–control studies conducted in Europe , but not in those conducted in North America  and Asia  (3 and ≥200 Bq/m3)  . The res0 Bq/m3) . Regardi2 < 50%). Statistically significant associations were observed in European studies and studies conducted for <10 years, with pooled ORs of 1.36  and 1.49 , respectively. Moreover, the adverse effect of a residential radon level of ≥200 Bq/m3 on childhood leukemia was confirmed. Notably, all subgroup analyses achieved high statistical significance without heterogeneity (I2 = 0%–42.35% and p ≥ 0.1).The subgroup analyses of studies of childhood leukemia were performed using fixed-effect models because of insignificant heterogeneity . With a 100 Bq/m3 increase in the residential radon level, the risk of lung cancer was not significantly different between the case and control participants . Therefore, a dose–response relationship was concluded to exist between all lung cancers and residential radon exposure.In this study, only 15 included studies provided relevant data regarding lung cancer risk due to a 100 Bq/m studies ,46,48,59Funnel plots were constructed to detect publication bias. The results indicated that articles with positive OR/RR values have been published, while articles with negative OR/RR values have not been published, leading to an unsymmetrical publication bias A,B. In c3 increase in the residential radon level, the lung cancer rates were not significantly different between the case and control patients.In this study, the causal associations between residential radon exposure and the increasing incidence of human cancers were assessed through a systematic review and meta-analysis of case–control studies. According to the eligibility criteria, 55 case–control studies with a total of 150,175 participants were enrolled and evaluated, and the results showed that radon exposure was strongly associated with increased lung cancer and childhood leukemia. However, with a 100 Bq/m2 = 42.28% < 50%), but not in North America and Asia. One out of 10 lung cancer cases in Europe is reportedly attributable to indoor radon exposure, and naturally occurring indoor residential radon may have caused around 19,000 lung cancer deaths in Europe in 2019  and 1.21 , respectively. Finally, the analysis of smoking status indicated a strong association between radon and lung cancer. The results revealed that in the same areas with high radon concentrations, smokers had higher lung cancer incidence than nonsmokers . Thus, it can be said that radon and smoking can be considered to have a synergistic effect on the occurrence and development of lung cancer.For assessing the relationship between radon exposure and lung cancer, 42 case–control studies that provided estimated size effect values (39 ORs and 3 RRs) were selected to perform a meta-analysis. However, the strong relationship was only confirmed through the meta-analysis based on OR values with a pooled OR of 1.36 , even though the analysis had high heterogeneity. The significant heterogeneity in the statistical analysis can be explained by the fact that these included studies were conducted in various regions of the world and under different radon exposure concentrations with different periods of investigation. To reduce the heterogeneity, a subgroup analysis was then designed to reevaluate the relationship based on subcriteria such as the study population, study period, and residential radon exposure. The results revealed a significant difference in lung cancer rate between the cases and controls in Europe  without heterogeneity (I in 2019 . The hig in 2019 . Specifi in 2019 . Owing t2 = 0% and p = 0.51). In contrast, the pooled RR of 1.15  (I2 = 52% and p = 0.08) from five case–control studies suggested no significant difference between childhood leukemia in the case and control groups. The controversial results may be due to the small size of the analysis, with eight studies reporting OR values versus five studies reporting RR values. Thus, adverse effects must be analyzed when larger sample sizes are obtained in the future. Similar to the lung cancer issues, the impact of radon on childhood leukemia was analyzed based on three subcriteria: study population, period of investigation, and radon level. The results also indicated that European children were more affected by residential radon than North American children, with pooled ORs of 1.36  and 1.89 , respectively. Investigations with a shorter duration (≤10 years) seemed to yield highly accurate results for predicting adverse effects compared with a longer duration (≥20 years). Notably, the impact of radon exposure was confirmed in participants exposed to high radon levels (≥200 Bq/m3) with OR of 1.56  (I2 = 42.35% and p = 0.188). Therefore, a high residential radon level over a long period is concluded to be positively associated with childhood leukemia; it mutates cells in the immune system and bone marrow, leading to DNA damage and influencing other biologic mechanisms, and consequently inducing childhood leukemia   determined the risk of lung cancer associated with radon exposure through a collaborative analysis of 13 case–control studies in nine European countries [ determinn Europe . Li et an Europe . Moon etn Europe . Poolingleukemia . All of The meta-analysis performed had three primary limitations. First, this study included only case–control studies; these may provide an incomplete overview of the causal association between residential radon and human health. Therefore, this topic should be addressed by analyzing all types of studies, such as cohort studies, ecological studies, and randomized clinical trials in the future. Second, the results may be affected by the classification of radon exposure. Given the different radon concentrations in other regions, the radon exposure classifications designed for each study are also different. Third, because it extracted only the OR/RR values of all lung cancers and all childhood leukemias, the study only investigated the overview of radon-induced cancers. Therefore, detailed conclusions about the incidence of each histological type of lung cancer  and childhood leukemia  could not be drawn.3) had a higher incidence of cancers compared to those in other regions . Thus, appropriate technical and policy solutions should be recommended to reduce radon exposure to ensure the health of environmental conditions and residents. Further, although the meta-analysis revealed the high impact of residential radon on human health, the results were limited by their high heterogeneity and unsymmetrical publication bias. Thus, this investigation should be repeated when more relevant articles are available in the future to overcome the current limitations.This meta-analysis of 55 case–control studies, including a total of 49,805 case patients and 100,370 control patients, suggested that residential radon was a risk factor for lung cancer and childhood leukemia when using estimated OR values for the analysis versus RR values. Notably, participants who lived in Europe or an area with high radon levels (≥100 Bq/m


Sorting and assembly machinery component 50 homolog (SAMM50) gene single-nucleotide polymorphisms (SNPs) have been connected with the susceptibility of nonalcoholic fatty liver disease (NAFLD), but with inconsistent results across the current evidence. The present work was schemed to explore the association between SAMM50 gene SNPs and NAFLD vulnerability via meta-analysis.PubMed, Web of Science, Cochrane Library, China National Knowledge Infrastructure (CNKI), and Wanfang were retrieved for eligible literature previous to June 10, 2021. The odds ratios (ORs) of the dichotomic variables and the standardized mean difference of quantitative variables with corresponding 95% confidence intervals (95% CIs) were computed to evaluate the strength of the associations. The quality of included studies was assessed using Newcastle-Ottawa Scale (NOS).rs2143571, rs3761472, and rs738491 polymorphisms respectively. The pooled data revealed that the 3 polymorphisms had conspicuous associations with NAFLD susceptibility: rs2143571, A vs. G, OR=1.51, 95% CI, 1.37–1.66, P < .01; rs3761472, A vs. G, OR=1.50, 95% CI, 1.35–1.67, P < .01; rs738491, A vs. G, OR=1.51, 95% CI, 1.40–1.63, P < .01.In total, 8 case-control studies encompassing 6297 NAFLD patients and 7306 disease-free controls in this meta-analysis. Ultimately, this analysis included 8, 6, and 5 studies for rs2143571, rs3761472, and rs738491 polymorphisms of the SAMM50 gene are appreciably associated with augmented risk of NAFLD vulnerability. It will provide the latest evidence to support the susceptibility of SAMM50 gene polymorphisms and NAFLD, and provide strategies for the prevention and treatment of NAFLD.This meta-analysis suggests that  It is gradually considered the liver disease component of metabolic syndrome, which is a risk factor for further development of fatty liver, nonalcoholic steatohepatitis (NASH), fibrosis, cirrhosis, and hepatocellular carcinoma.–4 The global burden of NAFLD is rapidly increasing and is expected to become the most common indication for liver transplantation with the alarming rise in the prevalence of obesity and metabolic syndrome. According to the World Gastroenterology Organization, NAFLD has been increasing during the past 20 years and is now one of the most common types of liver disease in Western countries. According to the present data, NAFLD affects 10%–30% of the general population in various countries, which has been viewed as a huge global health burden.,8Nonalcoholic fatty liver disease (NAFLD) is an assessment of fatty infiltration in the liver by histopathologic examination in the absence of excessive alcohol intake, autoimmune, viral infection, or drug-induced liver disease. At present, genome-wide association studies (GWAS) have demonstrated several conspicuous genetic susceptibility genes such as PNPLA3, SAMM50, PARVB, APOC3, PPAR-γ, NCANCILP2, FABP1, AGTR1, GSTs, which had been verified the crucial roles in the disease onset and progression of NAFLD.–15 Among them, rs738491, rs2143571, and rs3761472 in the sorting and assembly machinery component 50 homolog (SAMM50) gene were demonstrated to be notably associated with NAFLD susceptibility. Up to now, there have been several reports on SAMM50 gene SNPs and NAFLD susceptibility, but the results were inconsistent. Additionally, the real situation of the whole population cannot be reflected due to the limited sample size and regional and ethnic differences in the existing single studies. Therefore, a meta-analysis will be conducted on the existing research results to obtain more authentic and reliable results.Little is known about the latent mechanism involved in the development and pathogenesis of NAFLD, and yet it is a complicated metabolic process in which both environmental and genetic factors are etiology.This study depended on previously published literature and public databases. Thus, this work does not need ethical approval and patient consent. Two independent researchers searched the PubMed, Web of Science, Cochrane Library, CNKI, and Wanfang databases before June 10, 2021. The searching strategy of PubMed was exhibited as follows:  AND (“Sorting and assembly machinery component 50 homolog” [Mesh] OR SAMM50) AND (“Single Nucleotide Polymorphism” [Mesh] OR Variant OR SNP OR Polymorphism OR Mutant OR Mutation OR Variation). No language restriction was set. This study depended on previously published literature and public databases. Thus, this work does not need ethical approval and patient consent.Original studies were incorporated into this analysis in the light of the following inclusion criteria: (i) case-control studies; (ii) study investigated the associations of SAMM50 polymorphisms and NAFLD predisposition; (iii) NAFLD is diagnosed by pathology or ultrasound; (iv) control subjects were disease-free individuals; (v) detailed genotype data can be calculated for ORs and 95% CIs. Correspondingly, the exclusion criteria were as follows: reviews, conference abstracts, commentary articles, letters to the editor, animal studies, unpublished data, case reports, and secondary causes of steatosis, including alcoholism, total parenteral nutrition, hepatitis B/C virus infection, and drug usage.The Newcastle–Ottawa Scale (NOS) was performed to evaluate the methodological quality of included studies by 2 independent investigators. The NOS is composed of 3 aspects, namely selection, comparability, and exposure. Each study could receive 0 to 9 scores. Nevertheless, studies with ≥6 scores were regarded as high-quality studies. Disagreement was resolved by discussion.The extracted information contained the name of the first author, year of publication, country, ethnicity, numbers of NAFLD and control subjects, diagnostic methods of NAFLD, genotype frequencies, and Hardy–Weinberg equilibrium (HWE) results. Two investigators extracted the information independently. Summary ORs with 95% CIs were computed to specifically evaluate the relationship between SAMM50 gene polymorphisms and NAFLD susceptibility. Cochran’s Q-test and I2 test were applied to appraise the between-study heterogeneity. The random-effect model was employed to calculate merged ORs if P < .1, I2 > 50%. If not, the fixed-effect model was utilized for data synthesis. Sensitivity analysis was performed as well to assess the stability of all incorporated studies via the leave-one-out method. When the included studies were more than 10, a funnel plot was employed to evaluate the publication bias. On the contrary, less than 10 studies are not required. All analyses were done using Review Manager 5.3 software .In the case of polymorphism with 2 alleles (A1 and A2), both the NAFLD and disease-free participants could be classified into 4 exposure groups . Because the underlying genetic model for SAMM50 gene polymorphisms and NAFLD was unclear, the estimation of the association was based on 5 genetic models including allele contrast (A1 vs. A2), homozygous contrast (A1A1 vs. A2A2), heterozygous contrast (A1A2 vs. A2A2), dominant contrast (A1A1 + A1A2 vs. A2A2), and recessive contrast (A1A1 vs. A1A2 + A2A2). The control participants of incorporated studies were estimated via HWE.A primary search of electronic databases retrieved 26 potentially relevant publications: 13 from PubMed, 12 from Web of Science, 0 from Cochrane Library, 0 from CNKI, and 1 from Wanfang. No additional records were acquired from other sources. And then 20 studies remained after removing duplicated articles. Subsequently, a total of 4 irrelevant articles were excluded based on titles and abstracts screening. After applying inclusion and exclusion criteria, 3 unrelated articles (not related to NAFLD/SAMM50 gene polymorphisms), and 3 reviews were removed and 2 full-text article was not available. Ultimately, 8 studies  went into the process of meta-analysis. Overall, a flowchart summarizing the procedure of literature identification was illustrated in Figure –23 India, China, USA, Mexico, and Korea. They were all case-control designed and published between 2012 and 2018. It is worth noting that 2 studies  were genome-wide association studies (GWAS), including 2 stages of discovery I and validation II. In total, 8 studies encompassing 6297 cases and 7306 control participants were analyzed in the current analysis. For rs2143571, rs3761472, and rs738491 polymorphisms, there were 8, 6, and 5 studies ultimately incorporated, respectively. The main characteristics of the included literature were shown in Table  and Kanth et al did not provide the genotype frequency of the control group, so HWE could not be calculated. Studies with≥6 scores were regarded as high-quality studies according to the evaluation of methodological quality , homozygous model , heterozygous model , dominant model , and recessive model   including 6297 NAFLD patients and 7306 controls for rs3761472 to perform data analysis. The allelic model was analyzed by a random-effect model owing to significant heterogeneity observed. As the heterogeneity of genetic models was not significant, the fixed-effect model was employed for data analysis. The pooled data indicated there was a noticeable association between rs3761472 polymorphism and NAFLD susceptibility under allelic model , homozygous model , heterozygous model , dominant model , and recessive model  , homozygous model , heterozygous model , dominant model , and recessive model  (Table A total of 5 studies with 2150 cases and 4792 controls were used for data pooled. The fixed-effect model was employed for data analysis owing of the between-study heterogeneity was not significant. The results indicated an obvious association between P value, ORs, and 95% CIs did not change substantially. Therefore, the outcomes were considered to be statistically robust and reliable. The funnel plot for assessment of publication bias was not implemented on account of fewer than 10 research.After the omission of an individual study, the recalculated  In addition, SNPs in humans might be one of the critical steps to disclosing the genetic factor for NAFLD pathogenesis. In recent years, researchers have found that many gene polymorphisms are associated with NAFLD vulnerability through GWAS. Kawaguchi et al. were the first to find a significant association between SAMM50 gene polymorphisms and NAFLD susceptibility in the Japanese population via GWAS. Subsequently, the study was carried out by researchers from different countries, but the results were inconsistent. This inconsistency might be caused by factors like limited sample sizes, confounding factors, as well as clinical heterogeneity of NAFLD. Therefore, we collected the existing evidence and looked into the associations of SAMM50 gene SNPs and NAFLD vulnerability via meta-analysis, which could combine data from individual studies, examine and explain the heterogeneity, and increase the statistical power. In conclusion, the merged data suggested a significant correlation between rs2143571, rs3761472, and rs738491 polymorphisms of the SAMM50 gene and NAFLD vulnerability. Two large samples of GWAS were included in the original studies, with sufficient assurance to ensure the reliability of the study results. Of note, rigid methodological quality assessment, heterogeneity, and sensitivity analysis also ensured the robustness of the study results.Despite the specific pathological mechanism of NAFLD still needs to be explored. Nevertheless, research increasingly revealed that genetic predisposition plays a crucial intrinsic role in the occurrence and development of NAFLD.rs2143571, rs3761472, and rs738491 in the SAMM50 gene have a higher risk of NAFLD. Namely, it has a significantly increased risk for NAFLD. Furthermore, the outcomes of the meta-analysis were considered to be statistically robust and reliable according to the sensitivity analysis.In a word, we performed a meta-analysis of 8 case-control studies that satisfied the inclusion criteria. It should be noted that the current comprehensive analysis was more necessary and meaningful owing to the conclusions of qualified case-control studies are conflicting and contradictory. This work demonstrated that carriers of the A-allele of  It plays an important role in the formation of the mitochondrial crest, the assembly of the respiratory chain complex, and the maintenance of mitochondrial DNA stability. Studies have revealed that Sam50 reduction may lead to changes in mitochondrial shape and cristal morphology, and SAM50 consumption could affect the mitochondrial respiratory complexes. Ma et al. found a significant deletion of mitochondrial cristae in liver cells of patients with nonalcoholic steatohepatitis. In addition, changes in the normal structure of mitochondria could lead to insulin resistance, which is an important pathophysiological process of NAFLD.,35 All the above studies have indicated that the SAMM50 gene and its expression products play a crucial role in the development of NAFLD. According to the results of this meta-analysis, A allele of rs2143571, rs3761472, and rs738491 in the SAMM50 gene might be related to mitochondrial dysfunction and insulin resistance, leading to the NAFLD development.Sam50 is a kind of β-barrel protein distributed in the mitochondrial membrane and encoded by the SAMM50 gene.Up to now, this is the first synthetical study on the relationship between SAMM50 gene SNPs and NAFLD vulnerability. There were several strengths in this study. First, to gather a maximum amount of relevant literature, a comprehensive search strategy was adopted to retrieve eligible studies in both English and Chinese databases. Besides, the methodological quality of studies was evaluated via NOS, which allowed for the judgment of the potential risk of bias. According to the NOS, all eligible studies were of high methodological quality. Furthermore, sensitivity analyses were carried out in this study, which guaranteed the reliability of the findings.rs2143571, rs3761472, and rs738491 polymorphisms of the SAMM50 gene and NAFLD vulnerability.Nevertheless, there still existed several drawbacks that should be acknowledged. First, Although liver biopsy was regarded as the gold standard for NAFLD diagnosis, not all studies were included liver biopsy because pathology was expensive and invasive. Second, the original studies included in this study were mainly conducted in Asian and North American populations, and the results cannot reflect the situation of other populations. Third, We ignored the synergistic effect of polymorphisms at other sites of NAFLD because only 3 loci in the SAMM50 gene were studied in association with susceptibility to NAFLD. Thus, interactions between these loci and genes may result in concealing or amplifying the actual function of individual loci or genes. Leave aside these drawbacks, this study is the first to provide more accurate and powerful evidence on the association between rs2143571, rs3761472, and rs738491 polymorphisms of the SAMM50 gene are appreciably associated with an augmented risk of NAFLD vulnerability. In summary, this study will provide the latest evidence to support the susceptibility of SAMM50 gene polymorphisms and NAFLD, and provide strategies for the prevention and treatment of NAFLD. Concerning the limitations of this study, it is necessary to confirm the present findings with complementary studies with a larger sample size.This meta-analysis suggests that The authors would like to thank all studies and their participants involved in this investigation.Conceptualization: Ming Qiao.Supervision: Jian-hua Yang, Jun-ping Hu.Validation: Yi Zhu.Writing–original draft: Ming Qiao.Writing–review & editing: Jian-hua Yang, Jun-ping Hu.


Metabolic changes may occur following gastric surgery, which has been reported to contribute to bone loss, osteoporosis and even bone fracture. However, the evidence regarding the relationship between gastric surgery for benign and malignant conditions and risk of fracture is controversial. This study was conducted with the aim to evaluate whether gastric surgery is associated with a high risk of fracture.Major electronic databases were searched from inception through October 2021 for population-based cohort studies investigating the associations between gastric surgery  and risk of fracture compared with controls. Pooled relative risks (RRs) with 95% confidence intervals (CIs) were derived using the random-effects Mantel–Haenszel model. Multiple subgroup analyses and sensitivity analyses were carried out to test sources of heterogeneity stratified by various study characteristics and the robustness of the results.2 = 95.8%; P < 0.001] compared with that in control populations, among which the fracture sites of upper limb, spine, lower limb, pelvis and hip showed consistent significant results , whereas nonsignificant associations was noted for other fracture sites. Significant associations were also observed for patients having total or subtotal gastrectomy , gastric bypass , and a similar trend was observed for preserved passage procedures , though the difference did not reach statistically significant . An evident increased risk in the age range from 40-59 years was observed .A total of 14 studies comprising 693134 individuals were identified for analysis. The RR for the risk of fracture in people undergoing gastric surgery was 1.45 [95% confidence interval (CI) 1.23 - 1.72; IFrom this large pooled analysis of population-based cohort studies, evidence supports that fracture risk is increased in gastric surgery survivors compared with the control population. Early prevention and effective intervention strategies of bone fracture should be taken from clinicians and health policy makers.https://www.crd.york.ac.uk/prospero/display_record.php?RecordID=291394), identifier CRD42021291394PROSPERO ( People previously undergoing gastric surgery are subsequently at higher risk of fracture than control individuals.The highest fracture risk was seen after total or subtotal gastrectomy and gastric bypass.Increased risk for fractures was seen in the upper limb, spine, lower limb, pelvis and hip.Early prevention and effective intervention of bone fracture should be taken from clinicians in gastric surgery survivors.Bone fracture is a major public health problem worldwide, which causes a heavy economic burden and seriously affects the quality of life of the middle-aged and elderly adults. With the high rate of disability, fracture is also a major cause of premature death , 2. AlthGastric cancer, as one of the top burdensome cancers globally, represents the second commonest cause of cancer death globally. Surgical treatment remains the cornerstone of cancer cure and palliation . In addiThe inconsistent results of these studies prompted us to comprehensively assess the associations between gastric surgery and subsequent fracture risk through a systematic review. Moreover, we tried to explore the moderators, including study design, sample size, geographical regions, patient age, control population, fracture site, risk of bias, measurement of association, adjusted variables, and surgery type.https://www.crd.york.ac.uk/prospero/display_record.php?RecordID=291394).This study is reported in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA)  and MetaWe developed the search strategies for PubMed, EMBASE, and Cochrane Central Register of Controlled Trials without language restriction for original peer-reviewed articles published before October 31, 2021 investigating the associations between gastric surgery  and risk of fracture compared with controls. Terms related to the three primary concepts  were searched both as MeSH (Pubmed/Cochrane) terms or Emtree (Embase) terms and as text words. Full details for the complete search strategies and the search terms are provided in Studies were deemed appropriate for entry into the meta-analysis if they met the following inclusion or exclusion criteria: (1) study design: prospective/retrospective population-based cohort study; (2) participant: individuals previously underwent gastric surgery including bariatric surgery, total or subtotal gastrectomy for gastric lesions; (3) control: general populations having no history of gastric surgery matched or unmatched by demographic characteristics (4) the measure of association: studies reporting estimates including relative risk (RR), hazard ratio (HR), standardized incidence ratio (SIR) or incidence rate ratio (IRR) with corresponding 95% CIs that could be converted to the risk ratios. Cross-sectional studies, hospital-based or community-based observational studies and those providing inadequate data to generate precise estimates of the association between gastric cancer surgery and risk of fracture were all excluded. In addition, studies reported outcome of fracture resulting from metastatic cancer with bone localization, malnutrition/cachexia and bone loss due to the primary cancer were also excluded.Three authors compiled a piloted data extraction template and independently extracted data from each included study. In case of any discrepancies, discussion was initiated or the opinion of a senior author was requested. Several fields of general data were then extracted from each paper and entered into the data extraction template: first author of the study, publication year, study design, geographical region, study period, observation period, population sample size, participants’ mean or median age, control population, method of diagnosis of the cohort, outcome ascertainment, the main result of the study and measure of associations.For observational cohort studies, we used items from the Newcastle-Ottawa Scale (NOS) to evaluate methodological quality , with th2 statistic was used to quantify the amount of between-study heterogeneity . The pooled RR of fracture for people in the gastric surgery survivors compared with those in the general population or nonsurgery controls was the primary outcome measure. To account for the anticipated heterogeneity across studies, we employed the DerSimonian and Laird random effects meta-analysis to synthesize results (RRs with their corresponding 95% CIs) . Becauseogeneity . To furtogeneity . If publogeneity . SensitiOur literature search identified 491 eligible citations, 120 of which were excluded due to duplication. A further 308 were subsequently excluded based on title and abstract review, yielding 63 citations for full-text review. Because of a lack of outcome data, non-population-based cohort study design, reviews and meta-analyses without original data, a total of 14 studies comprising 693134 individuals satisfied the inclusion criteria and were eligible to be included in the final meta-analysis and quality assessment Figure 1Of the studies published between 1999 and 2021, 4 studies were conducted in the United States , 37, 39,The overall risk of bias was moderate to high for all studies based on the NOS tool. Bias was frequently seen in term of adequacy of follow-up followed by selection of control cohort and comparability of cohorts. We found that two studies , 35 was 2 = 95.8%; P< 0.001)  in survivors following gastric surgery compared with control populations. We noted significant inter-study heterogeneity (I< 0.001) Figure 2Subgroup analyses indicated that gastric surgery was associated with an increased risk of fracture among survivors with an age range from 40 to 59 years , while not among survivors ≥ 65 years  or with an age range from 30 to 39 years  Table 3.2 = 97.9%) and Asia (I2 = 94.6%), but was not detected in Europe (I2 = 0.0%). Similarly, heterogeneity was also significant in the subgroups of elder age (≥50 years)  and younger individuals (30-39 years) (I2 = 78.9%), fracture site of skull/face (I2 = 85.0%), upper limb (I2 = 88.8%) and lower limb(I2 = 86.2%), and surgery type of gastric bypass (I2 = 91.0%) and adjustable gastric banding (I2 = 84.1%), but was slight or not detected in the subgroups of fracture site of spine (I2 = 26.6%), pelvis and hip (I2 = 29.4%), the subgroup with surgery type of sleeve gastrectomy (38.1%). These analyses indicated that geographic region, age, surgery type, and fracture site could be potential sources of heterogeneity. Moreover, the residual heterogeneity could originate from other variation in demographic variables among the individuals of each included study.In addition, heterogeneity was high in the analysis of subgroups of studies conducted in the United States (IP= 0.381 for Begg’s test and P= 0.764 for Egger’s test).The trim and fill method used to adjusted for publication bias did not lead to imputation of any hypothetical missing studies, and the risk estimate remained the same.Using the leave-one-out sensitivity analyses, we further tested the stability of the result and indicated that no single study substantially altered the pooled risk estimates  . Early prevention and timely interventions are of great clinical significance in the prevention and treatment of these high-risk individuals. Moreover, the increased risk of fracture should be also mentioned during the preoperative informed consent process. Additionally, there is need for better understanding of the pathophysiological mechanisms, basic research on hormonal and neuro-intestinal pathways responsible for decreased bone quality.In conclusion, this meta-analysis showed that individuals who underwent gastric surgery may have an increased risk of fracture. Based on the subgroup analysis results stratified by most baseline variables, it is found that the results are still consistent and biologically plausible. However, before we get a high level of evidence based on prospective large cohort studies to prove this relationship, we should still interpret the results very carefully.The datasets presented in this study can be found in online repositories. The names of the repository/repositories and accession number(s) can be found in the article/Study concept and design: ZM, WZ, ZJ, ZC. Acquisition of data: QZ, CW, ZS, HW, ZM. Analysis and interpretation of data: CW, ZX, HW, ZM, WZ, ZJ, LC. Drafting of the manuscript: QZ, ZM. Critical revision of the manuscript for important intellectual content: All authors. Corresponding authors: ZM, WZ, ZJ, ZC.This study was supported by the Department of Science and Technology of Guangdong Province (No.2021A151511077) and the Traditional Chinese Medicine Bureau of Guangdong Province (No.20225012), and the Special Project for Clinical Research of Guangdong Provincial Hospital of Chinese Medicine (YN10101902) , the Guangdong Administration for Market Regulation  (No.〔2022〕158), the National Regional Traditional Chinese Medicine  Clinic Construction ((2018)205), Scientific Research Project from the Education Department of Fujian Province(No.JAT190205) and Joint Funds for the Innovation of Science and Technology, Fujian Province (No.2020Y9112).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.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. 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