Previous randomized controlled trials (RCTs) and meta-analyses of RCTs evaluating vitamin D supplementation for the prevention of cancer incidence and mortality have found inconsistent results and no meta-analysis has assessed the quality of the evidence available. We, therefore, aimed to perform an updated meta-analysis by including recent large-scale RCTs and assessing the quality of the pooled evidence.We searched several databases and trial registers from inception to April 2022. We used a random-effects model to estimate pooled risk ratios (RRs) and 95% confidence intervals (CIs). We used the Grades of Recommendation, Assessment, Development, and Evaluation (GRADE) considerations to evaluate the certainty of evidence.I2= 0%), total cancer mortality  and total mortality . The overall quality of evidence was high for all outcomes.We included 13 RCTs in our study. Vitamin D supplementation had no effect on the risk of total cancer incidence  with 109,543 participants.•Vitamin D is ineffective in reducing total cancer incidence and mortality in largely vitamin D-replete older adults.•Future RCTs should focus on populations with a higher prevalence of vitamin D deficiency. Meta-analysis; Vitamin D supplements; Cancer incidence; Cancer mortality. SeveralA previous meta-analysis of cohort studies suggested that higher serum levels of 25(OH)D are associated with lower cancer incidence and cancer mortality . AdditioSeveral RCTs, including the D-Health Trial, have been published after the most recently published meta-analysis. Therefore, this meta-analysis aims to update the literature and address the inconsistencies regarding the role of vitamin D supplementation compared with placebo in the prevention of the onset of cancer and cancer mortality.2Cochrane Handbook for Systematic Reviews of Interventions (Supplementary Table 1) [Our meta-analysis conforms to the guidelines presented in the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) statement and the Table 1) , 14. Our2.1ClinicalTrials.gov. from inception till April 2022 for any RCTs that involved a comparison of vitamin D at any dose with a control group. We used ProQuest Dissertations and Theses Global (PQDT) and OpenGrey to identify relevant grey literature. Bibliographies of previous meta-analyses and included trials were also screened to identify further RCTs related to our topic. No filter was applied in terms of language or publication period of the included studies.We searched the Cochrane Central Register of Controlled Trials , MEDLINE (via PubMed), Embase (via Ovid), and The complete search strategy for each database can be found in Supplementary Table 2.2.2We included all RCTs, regardless of the study population, that evaluated the effect of vitamin D supplementation  compared with placebo on total cancer incidence or mortality. Study designs other than RCTs such as observational studies were excluded. We also excluded those RCTs in which the total number of outcome events was ≤10 or the length of follow-up was ≤1 year.2.3All identified records were imported to Mendeley Desktop 1.19.8  where two authors (AS and MF) working independently screened them, first on the basis of titles and abstracts and then on the basis of full texts in accordance with the predefined eligibility criteria. Any disagreements were resolved by a third author (HAC).Data from included studies were extracted by two authors (AUR and MMH) into a pre-piloted Excel sheet. A third author (HAC) rechecked the completed extraction sheet and resolved any discrepancies and ensured the accuracy of the data. The following data items were extracted: study characteristics including the year of publication, study location, and follow-up duration, population characteristics such as age and gender, interventions , comparators, and outcomes (primary and secondary outcomes).2.4The primary outcomes were total cancer incidence and mortality. The additional outcome was all-cause mortality.2.5The methodological quality of our included studies was assessed by two authors (OB and AS) working independently using the revised Cochrane Risk of Bias Tool for randomized controlled trials (RoB 2.0) . The aut2.62 test and I2 statistic were calculated for each group or subgroup analysis to evaluate and quantify the heterogeneity present. I2 values were interpreted according to the Cochrane Handbook for Systematic Reviews of Interventions. [2 test was deemed statistically significant when P < 0.10 [Statistical analysis was performed using Review Manager . DerSimonian and Laird random-effects models were used to perform meta-analyses. Outcomes were reported as relative risk (RR) along with 95% confidence intervals. The Chientions.  The Chi2P < 0.10 .P-value less than 0.10 indicated publication bias.Funnel plots were constructed to assess publication bias and Egger's test was used to check funnel plot asymmetry using Jamovi (version 1.8) . A P-val2.7P < 0.05 was deemed statistically significant.We conducted subgroup analyses, on our primary outcomes, on the basis of the regimen of vitamin D supplementation, baseline and attained circulating 25(OH)D levels, and the different population groups . For the test for subgroup differences, For the primary outcomes, we also conducted a sensitivity analysis by excluding trials that tested the combination of vitamin D and calcium against a placebo.2.8Two authors (AS and HAC) separately evaluated the certainty of the evidence by using the Grades of Recommendation, Assessment, Development, and Evaluation (GRADE) tool which is based on five domains  , 18. We 33.1A comprehensive database search yielded a total of 3867 records while 3 articles were retrieved via citation searching. After removing duplicates (n = 983), the remaining articles were subjected to screening, leaving behind 14 reports of 13 RCTs, comprising 109,543 participants and spanning from 2003 to 2022, that were deemed eligible to be included in this meta-analysis  11, 12,, 29, 30.3.2Overall, two studies were found to have some concerns of bias, one in the domain of randomization and the other in the domains of randomization and deviations from intended interventions , 28, and3.33.3.13.3.1.1P = 0.56). There was no heterogeneity reported . The overall quality of evidence was rated as high (Supplementary Table 3). There was no statistically significant difference between any of the subgroups: the regimen of vitamin D intake (Pinteraction = 0.38), attained 25(OH)D level (Pinteraction = 0.64), baseline 25(OH)D level (Pinteraction = 0.55), and type of study population (Pinteraction = 0.47)  after excluding two trials that reported a combined effect of vitamin D and calcium against a placebo [Twelve RCTs were included in the analysis of total cancer incidence. The summary RR for intervention versus control group was 0.99  . The sen placebo .Figure 23.3.1.2P = 0.16). There was slight heterogeneity reported . The certainty of the evidence was rated as high (Supplementary Table 3). The sensitivity analysis reported an RR of 0.94  after excluding one trial that reported a combined effect of vitamin D and calcium against a placebo [Pinteraction = 0.25). Vitamin D decreased the risk of cancer mortality in participants with low baseline 25(OH)D levels (≤50 nmol/L) but not in those with baseline 25(OH)D levels >50 nmol/L; however, the test for subgroup differences was not significant ; daily intake of vitamin D was found to reduce cancer mortality  but no statistically significant association was observed between infrequent intake of regimen in the form of bolus and cancer mortality . There was moderate heterogeneity reported . No funnel plot asymmetry was detected with Egger’s test . The overall certainty of the evidence was rated as high (Supplementary Table 3).The summary RR for intervention versus control group was 0.92 .To the best of our knowledge, this is the most comprehensive meta-analysis to date evaluating the effect of vitamin D supplementation on cancer incidence and mortality. We found no significant association between vitamin D supplementation and total cancer incidence, cancer mortality and all-cause mortality. The overall certainty of the evidence was high for all outcomes suggesting that further research is very unlikely to change our confidence in the estimate of effect in largely vitamin D-replete older adult populations. The results also remained consistent across the subgroup analyses. Daily vitamin D intake was found to be associated with reduced cancer mortality compared to bolus intake; however, this observation could be due to chance as the et al. reported an inverse association between circulating levels of circulating 25(OH)D and total vitamin D intake with the incidence of colorectal adenoma and recurrent adenomas. However, the analysis did not find any association with supplemental vitamin D indicating that the supplemental sources of vitamin D may not be effective [Regarding total cancer incidence, our results are coherent with those reported by recent meta-analyses, which also showed no benefit of vitamin D supplementation in preventing cancer incidence , 31. Theffective . Meta-anffective . Resultsffective .More importantly, our findings disagree with those reported by recent meta-analyses, which suggested that vitamin D supplementation reduced cancer mortality , 31. Howet al., which showed a 7% reduction in mortality [We also found no reduction in all-cause mortality with vitamin D supplementation, which is congruent with a recent comprehensive meta-analysis  but contortality . Our updThe strengths of our study include it being the first meta-analysis to use the GRADE certainty of evidence considerations to inform its conclusions as well as being the largest meta-analysis to date, therefore, having a significantly increased statistical power compared to prior meta-analyses. We also assessed the effects of treatment on cancer mortality and incidence over a period of 3–10 years and excluded studies with few numbers of outcomes to increase the reliability of our results. Moreover, due to the variation in the designs of the studies, we were able to assess several pre-specified subgroups of interest. We did not investigate a large number of subgroups as the rate of type I error increases rapidly with the number of subgroup comparisons . We alsoWe would like to acknowledge a few limitations, however. First, there were a limited number of RCTs available and most of the included trials did not have cancer incidence and mortality as primary endpoints. Second, the majority of study participants consisted of Caucasians, and racial differences might modify the intervention effects . Third, interaction preclude any firm conclusions. Moreover, longer follow-ups may be required for cancers that progress over decades to demonstrate any benefit. Future large-scale RCTs should be conducted in populations with a higher prevalence of vitamin D deficiency to elicit any potential anticancer benefits of vitamin D supplementation.The null findings of our meta-analysis preclude any recommendations regarding the use of vitamin D supplements as a chemo-preventive measure. Importantly, however, most of the trials had few participants with a 25(OH)D concentration below 50 nmol/L, which is the population that is most likely to benefit. Our subgroup analysis suggested a reduction in cancer mortality in trials with baseline 25(OH)D less than 50 nmol/L; however, the scarcity of studies and a nonsignificant PIn conclusion, vitamin D supplementation is ineffective in reducing total cancer incidence and mortality in largely vitamin D-replete older adult populations. Future research should be based on populations with a higher prevalence of vitamin D deficiency and should involve more extended follow-up periods.Huzaifa Ahmad Cheema: Conceived and designed the experiments; Performed the experiments; Analyzed and interpreted the data; Contributed reagents, materials, analysis tools or data; Wrote the paper.Maurish Fatima; Anas Elgenidy: Performed the experiments; Analyzed and interpreted the data; Contributed reagents, materials, analysis tools or data; Wrote the paper.Abia Shahid: Conceived and designed the experiments; Analyzed and interpreted the data; Contributed reagents, materials, analysis tools or data; Wrote the paper.Oumnia Bouaddi; Mohammad Mehedi Hasan: Performed the experiments; Contributed reagents, materials, analysis tools or data; Wrote the paper.Aqeeb Ur Rehman; Salah Eddine Oussama Kacimi: Conceived and designed the experiments; Contributed reagents, materials, analysis tools or data; Wrote the paper.Ka Yiu Lee, PhD: Contributed reagents, materials, analysis tools or data; Wrote the paper.This research did not receive any specific grant from funding agencies in the public, commercial, or not-for-profit sectors.Data will be made available on request.The authors declare no conflict of interest.No additional information is available for this paper.


Recent investigations suggest that premature ventricular complexes (PVCs) during an exercise test are associated with an elevated risk of mortality in asymptomatic individuals. However, given the small number of studies included, the association between these two entities in the asymptomatic population remains obscure. Our aim was to evaluate this matter.A comprehensive literature search was conducted utilizing several online databases up to April 2022. The study comprised cohort studies examining the relationship between exercise-induced premature ventricular complexes (EI-PVCs) and all-cause mortality (ACM) as well as cardiovascular mortality (CVM) in asymptomatic populations. To provide diagnostic values across the statistically significant parameters, we additionally calculated sensitivity, specificity, and area under the curve (AUC).p < 0.001; I2 = 59.6%, p-heterogeneity < 0.001) and CVM (RR = 1.67 (95% CI = 1.40–1.99); p < 0.001; I2 = 7.5%, p-heterogeneity = 0.373). Subgroup analysis based on the frequency of PVCs revealed that frequent PVCs were similarly related to a higher risk of ACM and CVM, but not infrequent PVCs. Moreover, diagnostic test accuracy meta-analysis showed that recovery phase EI-PVCs have a higher overall specificity than exercise phase EI-PVCs regarding our outcomes of interest.A total of 13 studies consisting of 82,161 patients with a mean age of 49.3 years were included. EI-PVCs were linked to an increased risk of ACM (risk ratio (RR) = 1.30  = 1.18–1.42); EI-PVCs are correlated with a higher risk of ACM and CVM. When compared to the exercise phase, the specificity of PVCs generated during the recovery period in predicting interest outcomes is higher. As a result, we propose that the exercise ECG be utilized on a regular basis in middle-aged asymptomatic individuals to measure the frequency of PVCs and stratify the risk of mortality.https://www.crd.york.ac.uk/prospero/display_record.php?RecordID=328852], identifier [CRD42022328852].[ According to the 2019 European Society of Cardiology (ESC) guideline, an exercise electrocardiogram (ECG) is a valuable tool for confirming the diagnosis of coronary artery disease (CAD) in suspected patients, but it is not indicated in the asymptomatic population without suspicion toward CAD –5. AfterInterestingly, PVCs were also prevalent during exercise ECG in the asymptomatic group without a pre-existing CAD, with a frequency ranging from 3.8 to 27.5% , 9, 10. This meta-analysis was recorded in the PROSPERO  database under registration number CRD42022328852 and followed the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) standards.This systematic review and meta-analysis coveted to investigate whether EI-PVCs throughout exercise stress testing prognosticate mortality in both ACM and CVM in asymptomatic individuals. In favor of this aim, two independent investigators searched MEDLINE (via PubMed), Europe PMC, and ScienceDirect databases from inception to April 2022 with the search terms (“ventricular premature complexes” [MeSH Terms]) AND . We modified every search term to match the restrictions of each database, and the full-text articles were rigorously evaluated by both authors in case there was disagreement about whether research should be included based on its title or abstract.This study included both prospective and retrospective observational studies providing the occurrence of EI-PVCs in asymptomatic adult participants (age ≥ 18 years) who underwent exercise stress ECG testing, in conjunction with our primary outcomes of interest on clinically validated definitions of mortality, both prompted by any causes (ACM) and cardiovascular (CVM) origin. The studies must report the incidence of our primary outcomes in a comparative demeanor between patients with EI-PVCs to those without these arrhythmias (either during exercise or recovery phase of an exercise stress test). We excluded review articles, editorials, commentaries, case reports/series, meta-analyses, conference abstracts, and studies in languages other than English. The adjudicator would be the third investigator, and any discrepancies would be resolved through discussion.The preliminary search and duplication elimination were carried out by two separate investigators. Titles and abstracts were reviewed for eligibility, and the full text of potentially eligible papers was evaluated using predetermined inclusion and exclusion criteria. Data extraction from the suitable articles was conducted using a prebuilt table comprising the first author’s name, the origin of the study, study design, total subjects, and population of the study. For meta-regression purposes, several confounding factors such as age, sex, hypertension, diabetes mellitus, dyslipidemia, smoking status, and follow-up time were also included in the data gathering procedure.The risk of bias assessment for each and every study was performed independently by two authors using the Newcastle–Ottawa Scale (NOS) . A studyp-value <0.05. The I2 method was used to assess inter-study heterogeneity, and an I2 statistic >50% or p-value <0.10 indicated substantial heterogeneity.Software for Statistics and Data Science (STATA) software version 17.0 was utilized to conduct statistical analysis. This meta-analysis’ effect size was expressed as risk ratios (RRs) with 95% confidence intervals (CIs). Any studies that report hazard ratios (HRs) as their effect size were recast into RRs for the purposes of this analysis. The restricted maximum-likelihood method and random-effects models were used to pool the overall effect size despite their heterogeneity. All statistical analyses were two-sided, with statistical significance indicated by a Meta-regression for possible confounders mentioned in the previous subsection was also performed to ensure the true direct effect regarding the aforementioned exposure. Subgroup analyses were also conducted based on the timing and frequency of PVCs to produce an ideal, more robust summary across the relevant studies. The timing of PVCs was divided based on when the PVCs appeared throughout the exercise stress test (either exercise or recovery phase). Furthermore, frequent PVCs were classified as those occurring more than 5 beats per min or accounting for more than 10% of total ventricular depolarizations during the electrocardiographic recordings. We also employed diagnostic test accuracy (DTA) meta-analysis to generate diagnostic values across the statistically significant effect sizes and subgroups, which consisted of sensitivity, specificity, and area under the curve (AUC). Begg’s funnel plot and Egger’s test were used to assessing the likelihood of publication bias in both qualitative and quantitative terms, respectively.We identified 2,748 studies in our initial search, four of which were discovered via the hand-searching procedure, and 952 studies remained after duplicates were removed. Due to the fact that we used the minimum keywords in order to maximize the search results, the number of research dropped dramatically from 952 studies to 76 studies. During this pre-inclusion screening, the majority of the eliminated studies comprised PVCs that did not occur during the exercise stress test, pharmacological trials, diverse study populations, and animal studies. Conclusively, following a screening procedure based on pre-determined inclusion criteria, thirteen studies with a total of 82,161 participants were incorporated in this meta-analysis  7, 8, 1–22. The p < 0.001; I2 = 59.6%, p-heterogeneity < 0.001)  , as well as during recovery phase (RR = 1.46 (95% CI = 1.21–1.76); p < 0.001; I2 = 60.9%, p-heterogeneity = 0.026), are related with an elevated risk of ACM. Furthermore, subgroup analyses in the CVM outcome show a substantial increase in the risk of CVM in those who had EI-PVCs during the exercise phase (RR = 1.76 (95% CI = 1.32–2.35); p < 0.001; I2 = 35.3%, p-heterogeneity = 0.172) as well as people who had EI-PVCs during the recovery period (RR = 1.53 (95% CI = 1.18–1.98); p = 0.001; I2 = 0.0%, p-heterogeneity = 0.684).In this meta-analysis, we observed that patients who exhibit EI-PVCs during an exercise stress test have a greater risk of ACM (RR = 1.30 (95% CI = 1.18–1.42); < 0.001)  and CVM = 0.373) . The stap < 0.001; I2 = 53.1%, p-heterogeneity = 0.006), and (RR = 1.70 (95% CI = 1.42–2.03); p < 0.001; I2 = 3.8%, p-heterogeneity = 0.403), respectively, whereas the overall RRs of ACM and CVM in relation to infrequent PVCs were (RR = 1.37 (95% CI = 0.96–1.95); p = 0.080; I2 = 72.0%, p-heterogeneity = 0.013), and (RR = 1.41 (95% CI = 0.62–3.20); p = 0.411; I2 = 55.1%, p-heterogeneity = 0.135), respectively, which did not reach statistical significance. Information on the specific subtypes of the PVCs  that occurred within the overall included studies was accessible. Thus, we also performed subgroup analyses based on the previously mentioned categorization, which revealed that the overall combined RRs of ACM and CVM regarding frequent PVCs were (RR = 1.27 (95% CI = 1.15–1.39); With reference to our previous findings that both exercise and recovery phase EI-PVCs were significantly associated with a higher risk of ACM and CVM, summary receiver operating characteristics (SROC) were drawn, and AUC was calculated to determine the diagnostic value of each phase in the exercise stress test. EI-PVCs in the exercise phase had a sensitivity of 46% 25–69%), specificity of 66% (48–80%), and AUC of 60% (56–64%) for ACM outcome. Moreover, it possesses a sensitivity of 16% (7–33%), specificity of 85% (64–95%), and AUC of 45% (41–49%) during the recovery phase. Besides, pertaining to CVM, it was revealed that EI-PVCs in the exercise phase have a sensitivity of 23% (8–52%), specificity of 85% (70–93%), and AUC of 66% (62–70%). Statistical analysis apropos to the recovery phase in CVM demonstrated a sensitivity of 19% (5–49%), specificity of 90% (54–98%), and AUC of 51% (47–55%). –69%, spep > 0.05).An additional meta-regression analysis was carried out to discover if there was a link between prospective covariates and outcomes of interest in this study. The findings revealed that EI-PVCs occurrence and mortality risk were not affected by age, sex, hypertension, diabetes mellitus, dyslipidemia, smoking status, and follow-up time  (The heterogeneity statistic  > 0.05) , 3.The major findings of this meta-analysis are as follows. Firstly, EI-PVCs increased the risk of ACM and CVM substantially. Secondly, only frequent PVCs are significantly correlated with an elevated risk of ACM and CVM. Thirdly, the specificity of PVCs induced during the recovery phase was higher compared to PVCs that occurred during the exercise phase. Fourthly, EI-PVCs were more significantly associated with ACM and CVM in the short follow-up period (<7.5 years).Several included cohort studies revealed that patients with EI-PVCs were male, older, smokers, and more likely to have several comorbidities such as hypertension, diabetes mellitus, and dyslipidemia , 22, 23.Various hypotheses have been postulated to explain how PVCs induced by exercise testing in an asymptomatic population might enhance the risk of mortality. For starters, the PVC itself can act as the source of ventricular tachyarrhythmia, which increases the risk of sudden cardiac death (SCD) . Second,As the adage goes “structural heart disease may be silent and go undetected,” it might be one of the foremost inimical risk factors for mortality, given the majority of the included studies did not complete the imaging test at the beginning of the studies , 20–22. In the asymptomatic population, PVCs are denoted as a consequence of increasing automaticity in dormant ventricular pacemakers or can be generated by cardiotoxic drugs  or electrolytes disturbances  . MeanwhiWhereas parasympathetic dysfunction through vagal activity impairment can occur in certain individuals during the recovery phase, creating a pro-arrhythmic substrate that leads to PVCs . A meta-This meta-analysis indicates that EI-PVCs have poor sensitivity but good specificity in accordance with the diagnostic test accuracy meta-analysis. Thus, it is better utilized to rule in rather than rule out the probability of the aforementioned outcomes of interest. However, it should be noted that the mean age in the majority of the included studies only varied from 43 to 59 years. Thereby, the association between EI-PVCs and mortality in other age groups remained equivocal. We strongly suggest that exercise ECG, as a convenient and widely available test, be performed as a routine auxiliary test in the middle-aged asymptomatic population in order to evaluate the frequency of PVCs and stratify the risk of mortality.This study offers various advantages over previous meta-analyses. Firstly, the majority of the included studies (10 of 13 studies) have prospective designs, that diminish the risk of recall and selection bias. Secondly, in comparison to the Kim et al. study, this study contained a sufficient number of studies to perform the analysis between PVCs during the recovery phase and interest outcomes, thus the results were able to answer the cause of the moderately high inter-study heterogeneity in the previous meta-analysis. Lastly, the fact that all of our included studies employed multivariable-adjusted models, specifically ischemic ECG changes, further reinforces the value of this study’s pooled risk estimates, demonstrating that EI-PVCs may be used as a prognosticator in predicting mortality.However, this study still had several major limitations. First, undetected structural heart disease at the baseline could still partly explain the link between EI-PVCs and mortality. Second, low-to-moderate heterogeneity in the ACM analysis may be attributed to variable criteria to classify PVCs across the included studies. Third, due to the majority of included studies enrolling middle-aged participants, the correlation between EI-PVCs and mortality in the younger and older populations remained unclear. Therefore, future prospective cohort studies consisting of age groups other than this range with fixed frequent PVCs criteria and without structural cardiac disease screened by imaging modalities are needed to address the aforementioned issues.This study demonstrated that EI-PVCs in both phases (exercise and recovery) substantially raised the risk of ACM and CVM. Nonetheless, only frequent PVCs were shown to be correlated with ACM and CVM. Furthermore, the specificity of PVCs elicited during the recovery phase in predicting the interest outcomes is superior compared to the PVCs in the exercise phase. Hence, we propose that the exercise ECG be employed as an adjunct test in the middle-aged asymptomatic populations to measure the frequency of PVCs and stratify the risk of mortality.The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.MIQ and ICSP conceived and designed the study. ICSP, WK, and RP performed study selection, data extraction, and interpreted the data. MIQ, ICSP, WK, and RP performed extensive search of relevant topics. ICSP, WK, and RP performed the statistical analysis. CA, GK, MP, HG, MRA, ASK, and YHK performed review and extensive editing of the manuscript. All authors contributed significantly to the writing of the manuscript and approved the final manuscript.


Increasing evidence suggests that metformin can affect bone metabolism beyond its hypoglycemic effects in diabetic patients. However, the effects of metformin on fracture risk in type 2 diabetes mellitus (T2DM) patients remain unclear. A systematic review and meta-analysis were performed in this study to evaluate the association between metformin application and fracture risk in T2DM patients based on previous studies published until June 2021.A systematic search was performed to collect publications on metformin application in T2DM patients based on PubMed, Embase, Cochran, and Web of Science databases. Meta-analysis was performed by using a random-effects model to estimate the summary relative risks (RRs) with 95% confidence intervals (CIs). Subgroup analyses based on cohort/case-control and ethnicity and sensitivity analyses were also performed.2 = 96.8%). Moreover, metformin use also demonstrated similar results in subgroup analyses of seven cohort studies and four case-control studies, respectively . Sensitivity analysis revealed that there was no publication bias.Eleven studies were included in the meta-analysis. Results demonstrated metformin use was not significantly associated with a decreased risk of fracture (RR, 0.91; 95% CI, 0.81–1.02; IThere was no significant correlation between fracture risk and metformin application in T2DM patients. Due to a limited number of existing studies, further research is needed to make a definite conclusion for clinical consensus. The In contrast to patients with type 1 diabetes, T2DM patients exhibited increased or normal bone mineral density in the clinic but with increased bone fragility and fracture risk –15. The Metformin, a biguanide antidiabetic drug, is considered the standard initial treatment for T2DM patients. It affects several aging-related processes, including bone deterioration, by suppressing cellular senescence and chronic inflammation and promoting autophagy , 28. PreNonetheless, whether metformin can reduce the risk of fractures remains unconfirmed and controversial. Previous studies have reported no significant correlation between fracture risk and metformin application  and no sIn this study, to determine whether metformin treatment could reduce fracture risk in T2DM, a comprehensive meta-analysis was performed on the fracture risk of T2DM patients receiving metformin administration; it included all previous reports up to June 2021.22.1This meta-analysis was performed according to the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) statement guideline for systematic reviews and meta-analyses , 44, and2.2Each title and abstract were reviewed to identify relevant papers. Full texts of the articles were reviewed if the abstract was deemed potentially relevant. Studies that met the following criteria were eligible for inclusion:(1) observational studies where metformin was the exposure variable and fractures were the main outcome variable or one of the outcome variables.(2) T2DM participants aged ≥18 years.(3) The odds ratio (OR), risk ratio (RR), and hazard ratio (HR) were reported as the effect size (ES).2.3The exclusion criteria excluded studies that:(1) Related to other drugs in combination treatment with metformin.(2) Excluded placebo diabetic control.(3) Included Type 1 diabetes patients.(4) Included single gender.A total of 1,031 publications were identified with the search strategy. Then, these studies were independently screened according to the inclusion and exclusion criteria. A total of 11 studies were eligible and included in the meta-analysis. Figure 12.4Two authors (WYN and YLM) independently conducted study screening and data extraction from the eligible literature. When disputes were encountered, they were resolved through discussion or assisted by the main investigator (LYZ). The following data were collected from all included studies: first author’s surname, publication year, study design, country, follow-up duration, mean age or age range of participants, gender, sample size, number of cases, outcome variables and fracture assessment method, the adjusted ORs, RRs, or HRs, and the corresponding 95% confidence intervals (CIs) Table 1.2.52 tests were performed to analyze the homogeneity of the included studies. For the Q test, statistical significance was set at p ≤ 0.05; for I2 statistics, the following critical points were specified to define the degree of heterogeneity: <25% (low heterogeneity), 25%–50% (moderate heterogeneity), 50%–75% (high heterogeneity), and >75% (severe heterogeneity).Stata software was used for meta-analysis . All reported ORs, RRs, HRs, and the 95% CIs for fracture risk were used to calculate the logarithmic RR and its standard error (SE). A random effects model was used to estimate the summary relative risks with 95% CIs. Q and IP values of < 0.05 were considered statistically significant.The subgroup analyses of cohort, case-control, and ethnicity were respectively performed on the included studies. In addition, sensitivity analysis was used to investigate the extent to which inferences might depend on a particular study or research group. Visual inspection of the funnel chart was used to assess publication bias. A formal statistical assessment of funnel plot asymmetry was performed using Egger’s regression asymmetry test. 33.1Among the 1,031 retrieved papers, 11 related to the application of metformin and the risk of fracture in T2DM patients and were included in this meta-analysis according to the inclusion/exclusion criteria Table 1.These studies collectively included 635,945 participants and were published between 2008 and 2021; they were conducted in various regions, including one study each in the United States , DenmarkRegarding the types of fractures, five studies included fractures at multiple sites, such as fractures of the proximal and distal upper and lower extremities, ribs and thorax, hip, and foot , 51, 53,3.2There were seven cohort studies and four case-control studies involved in the meta-analysis. Four studies revealed that metformin treatment reduced fracture risk, seven studies demonstrated that metformin had no significant effect associated with fracture risk, and no studies showed that metformin increased fracture risk.2 = 96.8%, p < 0.001).In this meta-analysis, we had 11 effect sizes obtained from 11 studies. The meta-analysis results are shown as forest plots Figure 2p = 0.99).In the sensitivity analysis, no single study significantly influenced the findings. No evidence of publication bias was found in this meta-analysis with Egger’s test evaluation  being included in this study, investigation of fractures in specific sites was not possible; therefore, our study used the same strategy as that of a previous report  that foc2 = 98.0%, p < 0.001 . Inter-study heterogeneity was significant; I2 = 0.0%, which demonstrated that the high heterogeneity across all 11 studies did not significantly affect the results in this meta-analysis. Therefore, the conclusion of this meta-analysis was robust and credible. The Egger’s test result was p=0.9301 and p>0.05, which indicated that the meta-analysis has no publication bias. A sensitivity analysis was used to test the stability of the effect size (ES) estimates  . The reduced fracture risk with metformin could be related to metformin prescriptions that typically start in the early stages of type 2 diabetes mellitus  . Anothermellitus . Furthermellitus .This study found that metformin use is not significantly associated with a decreased risk of fractures in T2DM patients. Our results are inconsistent with previous studies including type 1 and type 2 patients and single-gender data. Our study focused on T2DM patients and only included studies that examined both genders. A previous study by Wallander et al. , demonstIn summary, our data demonstrated that metformin treatment was not significantly associated with the risk of fracture, and our results are independent of patients’ HbA1c levels/glycemic control levels. When the data with HbA1c control is considered for analysis, for example, If the data of HbA1c ≤ 7.0%    was usedp-values for interaction<0.05). The protective effect of metformin was not significant in insulin-treated patients, while metformin revealed greater beneficial effects in sulfonylurea or pioglitazone-treated patients  on bone markers P1NP Procollagen type 1 N-terminal propeptide (P1NP) and C-terminal telopeptide of type I collagen (CTX) in patients with T2DM . The levpatients . The posThis meta-analysis has some limitations. There was significant heterogeneity between the 11 included studies. The reason may be due to differences in the sample sizes of the included studies. For instance, the study of Starup-Linde et al., 2017  included5In this systematic review and meta-analysis, we found that metformin administration was not significantly correlated with a decreased fracture risk in T2DM patients. These results were independent of patients’ HbA1c levels and glycemic control levels. Due to the limited number of studies included in this meta-analysis, further investigations are needed to make stronger conclusions for clinical consensus.The original contributions presented in the study are included in the article/Study design/manuscript preparation: YL, LL, and LC. Original draft preparation/Meta analysis: YW and LY. Sub-Group Meta analysis and validation: ZY and RL. Review and editing: AS, LL, and JW. Conduct data/Publication collection/Analysis: FD, and XZ. Funding acquisition: YL, LL, and LC. All authors have read and agreed to the published version of the manuscript. All authors contributed to the article and approved the submitted version.


With the growth and aging of population, the incidence of prostate cancer will increase year by year, which is bound to bring greater economic burden to the society. There has been greater interest in the anticancer effects of statin in recent years. It is controversial whether statin use is associated with the risk of prostate cancer (PCa). Thus, we conducted a meta-analysis and systematic review to explore the effects of statin use and their duration and cumulative dose on the overall incidence of PCa.  The study was conducted according to the latest guidelines for PRISMA 2020. We searched PubMed and other databases for studies about the association of statin use with the risk of incident prostate cancer between January 1, 1990, and April 11, 2022. Two independent researchers extracted data and evaluated the quality of the studies. R x64 4.1.2 and random-effects model were used for data statistics. Relative risk (RR) and odds ratio (OR) effective values with a 95% confidence interval (95% CI) were used to assess the main results. I2 = 98%, P < 0.01; case-control study: I2 = 72%, P < 0.01).  The results of 6 RCT and 26 cohort studies showed that statins did not significantly associate with the incidence of PCa . The similar results were obtained from 9 case-control studies . However, statins were associated with a lower risk of Pca  when the cumulative defined daily dose (cDDD) was high. Using statins for more than five years could be associated with a reduced incidence of Pca . There was a significant heterogeneity in these studies (RCT and cohort study:  We concluded that statins had a neutral association with the overall risk of PCa. High cDDD and long duration were associated with a lower risk of PCa. Prostate cancer (PCa) is the second most commonly diagnosed cancer and the sixth leading cause of cancer death in men, with an estimated 1.4 million diagnoses worldwide in 2020. With the growth and aging of the population, it is expected that the burden of PCa will grow to 2.3 million new cases and 740,000 deaths by 2040 [ABCA1 promoter and subsequent transcriptional silencing. In addition, a study [Statins are 3-hydroxy-3-methylglutaryl-CoA (HMG-CoA) reductase inhibitors, which are comprehensive lipid-regulating agents. It can not only effectively decrease total cholesterol (TC) and low-density lipoprotein (LDL) but also decrease triglyceride (TG) and increase high-density lipoprotein (HDL). The mechanism of statins is that the intracellular cholesterol synthesis is decreased by competitively inhibiting the endogenous cholesterol synthesis rate-limiting enzyme (HMG-CoA reductase) and blocking the intracellular hydroxyvalerate metabolism pathway. Statins have been shown to work through mevalonate-dependent and mevalonate-independent mechanisms . There h a study  has show a study . TherefoMore than 100 articles have reported that statin usage is associated with PCa, including the overall incidence of PCa and the risk of local or advanced PCa and so on. A previous meta-analysis  that incThus, it is controversial whether statin usage is associated with the risk of PCa in different populations. Furthermore, the previous meta-analysis cannot well claim the effects of statins on prostate cancer that they did not include enough studies or their result analysis was incomplete. This meta-analysis and systematic review are a revision and update of all the previous clinical studies, focusing more on and aiming to unravel the effects of different types of statins and their duration and cumulative dose on the overall incidence rate of PCa.This systematic review and meta-analysis were reported according to the latest guidelines for PRISMA 2020  with 95% confidence intervals (CIs). As for case-control studies, the participants number of case group and control group and the number of statin users and non-users in the two groups also need to be recorded, respectively, to get odds ratio (OR) with 95% confidence intervals (CIs).Study Design. It included randomized controlled trials (RCTs) and cohort study and case-control study.All studies or literature included in the meta-analysis and systematic review strictly met the following PICOS  criteria:Two independent researchers (M-YX and J-XS) searched PubMed, Embase-Ovid, Cochrane Library, and ClinicalTrias.gov for publications between January 1, 1990, and April 11, 2022. We also searched some grey literature in Google Academic, American Society of Clinical Oncology (ASCO) meeting summary, European Urology Society (EAU) meeting summary, and American Urology Society (AUA) meeting summary. We used keywords to search for “Statin(s)” or “Atorvastatin” or “Cerivastatin” or “Compactin” or “Fluvastatin” or “HMG-CoA” or “Lovastatin” or “Mevastatin” or “Pravastatin” or “Rosuvastatin” or “Rosvastatin” or “Simvastatin” and “PCa” or “Prostate Neoplasms” and so on. Supplementary Material Two independent researchers (M-YX and J-XS) extracted data using a data extraction sheet that included basic information and some research factors of interest . Basic iTwo researchers (M-YX and J-XS) independently assessed the quality of cohort and case-control studies using the Newcastle-Ottawa scale (NOS). The NOS scale which includes eight items consists of three parts . A score of 7 to 9 is defined as high quality, and a score less than 7 is defined as low quality.We adapted RoB2 Excel to evaluate the quality of RCTs. RoB2 set up five evaluation domains: the randomization process, deviations from intended interventions, missing outcome data, measurement of the outcome, and selection of the reported result. There are many different signal problems in each domain. Researchers need to make judgments and answer these questions objectively when evaluating the bias risk of an RCT. There are generally five answers to signal questions: yes (Y), probably yes (PY), probably no (PN), no (N), and no information (NI). According to the reviewers' answers to the signal questions, the bias risk in each domain can be divided into three levels: “low risk,” “certain risk,” and “high risk.” Any differences are resolved by re-evaluating the original article with the third researcher.Q and I2 statistics to test the heterogeneity. The random-effects model should be adopted due to expected heterogeneity among different study populations. In order to verify the publication bias of studies, we conducted Begg's and Egger's tests [In RCT and cohort studies, the estimated RR effect values with 95% CI were used to assess the potential association between statin usage and PCa risk, whereas the OR effect values were used in case-control studies. We used Cochrane's 's tests . We condP value <0.05 indicated statistical significance.We used the “meta” package in R x64 4.1.2 to perform meta-analysis , 16. AllWe have retrieved 1,250 publications from electronic databases and grey literature. We remained 731 after removing repetitive literature. After a preliminary reading of the title and abstract, we retained 110 articles. After screening according to strict criteria, 57 articles were excluded. Then, we read the full text of each record. After excluding 12 publications due to a lack of data about the incidence of PCa, 41 studies were finally included in this meta-analysis and systematic review .The characteristics of all studies included are shown in Supplementary As for the quality of the included studies, more details are presented in Tables I2 = 98%, P < 0.01, I2 = 72%, P < 0.01, 6 RCT studies and 26 cohort studies showed that statin use was not significantly associated with reducing the incidence of PCa . Similar results were obtained from the analysis 9 case-control studies . The forest plot  showed tP value for interaction between the two types was 0.18, indicating that the study design did not contribute to the heterogeneity of the literature whereas different countries were one of the sources of heterogeneity because the P value for interaction in the subgroup analysis of countries was <0.01. Studies conducted in Denmark  and Israel  had shown that statins could be associated with decreasing the incidence of PCa. The studies of Scotland  and Sweden  showed that statins were associated with a higher incidence of PCa. Given that P value for interaction in the subgroup analysis of statin type was 0.13, there was no significant difference in the incidence of PCa among different statins.In order to find the source of heterogeneity, we conducted a subgroup analysis . In the 2, race (the percentage of white), PSA level, GS, and population percentage of cDDD >600 were constructed. For RCT and cohort studies, we found that race (the percentage of whites) (P = 0.0430) significantly affected the effects of statins on the incidence rate of PCa (P = 0.9890), follow-up period (P = 0.8267), age (P = 0.8852), BMI (P = 0.9106), and cDDD <600 (P = 0.8718) (Supplementary P = 0.8254) and follow-up period (P = 0.8254).In addition, we used meta-regression analysis for some continuous variables to further find the covariates. The meta-regression models of publication year, follow-up time, age, BMI, population percentage of BMI <30 kg/me of PCa . There w4 studies recorded the cDDD for statins of each group of participants in detail , 40, 48.z = −0.16, P = 0.8712) and Egger's test  indicated that there was no publication bias [z = 0.83, P = 0.4042) and Egger's test .In order to test the stability of our results, we conducted a sensitivity analysis  or cumulion bias . For OR In recent decades, researchers have paid more attention to the anticancer progress of statins. Statins are not only fully explored in PCa  but alsoStatins work through noncholesterol-mediated or cholesterol-mediated pathways . StatinsIn this meta-analysis, we included a total of 41 studies on statin usage and the incidence of PCa, and we used RR and OR as effect values. RR was mainly used in randomized controlled trials or prospective cohort studies to study the effect of statin exposure on the outcome of PCa by designing follow-up studies . The OR 6 RCT studies did not show a significant association between statin use and PCa. These 6 studies explored the relationship between statin use, all-cause mortality, and specific cancer incidence in hypercholesterolemia patients. Cardiovascular events were the main outcome, so there were basic diseases in patients. 10 of 26 cohort studies reported that statins were associated with a lower incidence of PCa, which was in line with our expectations. Only one of nine case-control studies indicated that statins associated with a lower incidence rate of PCa . In addiMany of these studies relied on questionnaires to obtain exposure data and lacked screening adjustments . NeverthWe identified documented cDDD in three cohort studies, and we found that high cDDD (cDDD >600) is significantly associated with a reduced risk of PCa. However, cDDD was the effect value accumulated by the defined daily dose over time, so it was difficult to identify whether long-term duration or a large dose led to the reduction of PCa risk. Then, to explore the impact of duration on the risk of PCa, we recorded and analyzed the specific duration of each group. In 3 cohort studies and 4 case-control studies, some participants used statins for more than 5 years. By analyzing the total RR effect value of these cohort studies and the total OR effect value of these case-control studies, we found that when the duration was more than 5 years, the overall risk of PCa in cohort studies decreased significantly, and the overall risk of PCa in case-control studies decreased but not statistically significant. Long-term use of statins had a more stable effect on the anticancer effect in the body.It shows high heterogeneity in the overall analysis. In some subgroups, heterogeneity was decreased, indicating that research types and countries were important influencing factors. Researchers could dictate that participants use a certain type of statins in an RCT. However, in observational studies, the type of statins used by participants could not be intervened. In addition, the definition of statin use was different in different studies. Thus, statins type was also one of the sources of heterogeneity.Meta-regression showed that the proportion of whites was positively correlated with the risk of PCa. On the one hand, this was due to the different genetic susceptibilities of different races. On the other hand, Caucasians were more likely to be diagnosed with PCa because of different levels of medical care in different countries.Concerning confounding factors, we put forward some thoughts. There was a direct correlation between the malignant degree of PCa and PSA . The higFor the prevention of PCa, we not only pay attention to statins but it is more important that residents maintain good living habits. The goal of the World Cancer Research Fund/American Institute for Cancer Research (WCRF/AICR) Report is to improve health and reduce the global cancer burden through diet, nutrition, and physical activity , which iCompared with previous meta-analysis, ours included more qualified literature, and the sample size was greatly increased, making the conclusion more universal. Moreover, we not only explored the effects of statin usage on the overall risk of PCa but also further explored the association between cDDD, the duration of statins, and the risk of PCa.Some limitations should be noted. First, some data were incomplete in different studies. For example, the follow-up period was not recorded in 3 studies, 16 studies lacked age data, 32 studies lacked BMI data, 25 studies lacked race composition data, 33 studies did not record participants' PSA baseline value. And the type of statins in 16 studies was unknown. Second, due to the inclusion of too many studies, it was difficult to unify each study's design methods and the definition of statin exposure, so that the heterogeneity between studies was high. Third, it is difficult to accurately record the long follow-up period and other vague data. Therefore, it remained to be seen whether a certain type of statin and its dosage and duration of use affects the risk of PCa. For future research, more clinical trials and in vitro laboratory experiments are needed to further verify the role of statins in PCa.In summary, we conducted a meta-analysis to explore the effects of statin use on the overall incidence of PCa. We found that statins had a neutral effect on the overall risk of PCa, and high cDDD and long duration were associated with a lower risk of PCa, which may contribute to preventing PCa and decreasing its incidence.


Body composition as dynamic indices constantly changes in pregnancy. The use of body composition indices in the early stages of pregnancy has recently been considered. Therefore, the current meta-analysis study was conducted to investigate the relationship between body composition in the early stages of pregnancy and gestational diabetes.Q test of heterogeneity, and I2 index was used to quantify heterogeneity.Valid databases searched for papers published from 2010 to December 2021 were based on PRISMA guideline. Newcastle Ottawa was used to assess the quality of the studies. For all analyses, STATA 14.0 was used. Mean difference (MD) of anthropometric indices was calculated between the GDM and Non-GDM groups. Pooled MD was estimated by “Metan” command, and heterogeneity was defined using Cochran’s Finally, 29 studies with a sample size of 56438 met the criteria for entering the meta-analysis. Pooled MD of neck circumference, hip circumference, waist hip ratio, and visceral adipose tissue depth were, respectively, 1.00 cm (95% CI: 0.79 to 1.20) , 7.79 cm (95% CI: 2.27 to 13.31) , 0.03 (95% CI: 0.02 to 0.04) , and 7.74 cm (95% CI: 0.11 to 1.36) .Increased neck circumference, waist circumference, hip circumference, arm circumference, waist to hip ratio, visceral fat depth, subcutaneous fat depth, and short stature increased the possibility of developing gestational diabetes. These indices can accurately, cost-effectively, and affordably assess the occurrence of gestational diabetes, thus preventing many consequences with early detection of gestational diabetes. Gestational diabetes mellitus (GDM) is defined as carbohydrate intolerance with varying degrees that is first diagnosed in pregnancy . GDM usu2, family history of diabetes, history of a macrosomic infant weighing 4.5 kg, and race , OR ’GD’ [tiab], OR ’Gestational Diabetes Mellitus’ [tiab], OR ’GDM’ [tiab], OR ’pregnancy induced diabetes’[tiab]’Anthropometric indicators’ [tiab], ’Anthropometric indices’ [tiab], OR ’body size’[tiab], OR ’body composition’ [tiab] OR, ’Waist/Hip Ratio’ [tiab], OR ’WHR’ [tiab], OR ’ visceral fat mass’ [tiab], OR ’VFM’ [tiab], OR ’ Neck circumference’ [tiab], OR ’hip circumference’ [tiab], OR ’ waist circumference’ [tiab], OR ’ subcutaneous adipose tissue’ [tiab], OR ’ skeletal muscle mass percentage’ [tiab], ’total adipose tissue thickness’ [tiab], OR ’subcutaneous adipose tissue’[tiab], OR ’Subcutaneous fat thickness’ [tiab], OR ’visceral adipose tissue depth’ [tiab], OR ’skinfold thickness’ [tiab], OR ’mid upper arm circumference’ [tiab], OR ’subcutaneous fat thickness’ [tiab], OR ’fat mass percentage’ [tiab], OR ’fat mass index’ [tiab], OR ’muscle mass percentage’ [tiab], OR ’Skinfold Thickness’ [tiab]’Pregnancy’ [tiab], OR ’Pregnancies’ [tiab], OR ’Gestation’[tiab], OR ’early pregnancy’ [tiab]#1 AND #2#1 AND #2 AND #3We set our inclusion and exclusion criteria based on PICO criteria  .The initial search yielded 3523 results. The eligibility of these articles was independently evaluated by two authors, and disagreements were resolved by consensus. In the first stage, 2108 irrelevant or duplicate articles were excluded. After reviewing the titles and abstracts of the remaining articles, 918 more papers were excluded. In the evaluation of the full texts, 139 ineligible articles were excluded out of the remaining 180 articles. Finally, a total of 41 eligible articles were reviewed and 29 articles meets criteria to meta-analysis .Newcastle Ottawa scale (NOS) was used to measure the quality of studies. This scale is used to measure the quality of cohort and case control studies. The validity and reliability of this tool have been proven in various studies , 36.Two authors independently performed the study selection and validity assessment and resolved any disagreements by consulting a third researcher. Author, year, study design, geographic region, maternal age, diagnostic criteria of GDM, anthropometric indices, accompanying factors, results, and quality assessment scores were extracted from articles.All analyses were conducted with STATA 14.0 . For each study, mean value and standard deviation (SD) of anthropometric indices were extracted; if IQR was reported, we changed it to SD with IQR/1.35. Then, the mean difference (MD) of anthropometric indices was calculated between GDM and non-GDM group for each study. Then, standard error (SE) of MD was calculated for each study using the following formula:1,2 are variance values, and samples size in GDM and control groups, respectively. Then, pooled MD was calculated by “Metan” command [Q test of heterogeneity, and the I2 index was used to quantify heterogeneity. In accordance with Higgins classification approach, I2 values above 0.7 were considered as having high heterogeneity. To estimate the pooled MD for anthropometric indices, the fixed-effect model was used; when heterogeneity was greater than 0.7, the random effects model was used. The meta-regression analysis was used to examine the effect of publication year, age, sample size, and study design as factors affecting heterogeneity among studies. The “meta bias” command [Where,  command . Heterog command  was used command . In all Twenty-nine studies with a sample size of 56,438 met the meta-analysis inclusion criteria . Fig 1 s2 index was above 70% for all mentioned indices. The current study set to investigate the relationship between body composition and GDM as a systematic review and meta-analysis. The results indicate that anthropometric indices such as WC, NC, HC, WHR, VAT, SAT, Height, and MUAC are associated with GDM; an increase in the indices of WC, NC, HC, WHR, VAT, SAT, and MUAC increase developing GDM, also short stature increases the susceptibility to GDM.We investigated that VAT and SAT are associated with GDM. Alwash et al.(2021) found that all three obesity phenotypes were significantly associated with the risk of developing GDM. In addition, visceral obesity was a stronger risk factor for GDM than other obesity phenotypes . Yao et De Souza et al.(2015) determined the relationship between SAT depth, TAT depth, and VAT depth in the first trimester of pregnancy and the occurrence of GDM in mid-pregnancy. It was observed that increasing the depth of VAT and TAT independently of BMI could predict the risk of dysglycemia in later stages of pregnancy . SimilarThe results of the present study also showed WC, HC and WHR are associated with GDM. Various studies have shown an association between WC and WHR-based central obesity around the hip with the occurrence of GDM . HoweverKansu-Celik et al.(2018) observed a significant relationship between 50g GCT and WC, and SAT thickness. He showed that SAT predicts thickness greater than 16.75 mm GDM with a sensitivity of 71.7% and a specificity of 87.6% . In adulWe have shown that increased NC also leads to GDM. Hancerliogullari et al.(2020) also stated that NC in women with GDM are significantly higher  and NC iIn this study we reported that short stature increases the susceptibility to GDM. Height in adulthood is an indices of genetic, early and childhood factors and their interactions. Although the biological mechanism associated with adult height and GDM is unknown, several pathways have been suggested. For example, malnutrition of the fetus may lead to low birth weight, which is associated with shorter height in adulthood, and may also be associated with metabolic disorders in adulthood. Height has different variations in different populations , 85. In Body composition in pregnancy has a dynamic process; for example, changes in weight gain and free body adipose mass during pregnancy are clearly observed .Measuring maternal body composition during pregnancy is challenged by existing in-vivo measurement methods that cannot distinguish between maternal and fetal reserves  and lookThe composition of the mother’s body changes during pregnancy to support optimal fetal growth. In the first few months of pregnancy, changes in the composition of the mother’s body indicate the readiness of the female body for fetal growth. Especially, the uterine and breast tissue that makes up the mother unit grows and the blood volume increases. In late pregnancy, more pronounced growth of the embryonic unit  occurs along with the continued growth of maternal tissue and further increase in blood volume. At the time of delivery, the fetal unit accounts for approximately one-third of the total GWG .Accordingly, central obesity is associated with more obesity-related complications . In contBody composition indices such as WC, HC, WHR, AC, VAT, SAT, and height can relate more effectively and accurately to GDM. These available anthropometric indices can be used as a tool to assess the occurrence of GDM in an accessible, cost-effective, and high-precision manner.One of the limitations of the study is the difference in the critical values of the criteria used to diagnose GDM, which may affect the decision on the absence or occurrence of GDM based on different indices. In addition, studies conducted in different populations and races, which is a determining factor in body composition and can affect both body composition and the occurrence of GDM, have not been considered in the present study. Also, the small number of studies performed on some anthropometric indices is another limitation of the study, which makes it difficult to draw conclusions about such indices.S1 File(DOC)Click here for additional data file.S1 Checklist(DOC)Click here for additional data file.


To examine the effect of diet quality on the risk of gestational diabetes mellitus.This review included cohort and case-control studies reporting an association between diet quality and gestational diabetes mellitus. We searched PubMed, Cochrane Library, Web of Science, Embase, PsycINFO, CINAHL Complete, Chinese Periodical Full-text Database, China National Knowledge Infrastructure, Chinese Biomedical Literature Database, and China Wanfang Database for studies published from inception to November 18, 2022. The Newcastle-Ottawa Scale was used for quality assessment, and the overall quality of evidence was assessed using the GRADEpro GDT.A total of 19 studies  with 108,084 participants were included. We found that better higher diet quality before or during pregnancy reduced the risk of developing gestational diabetes mellitus, including a higher Mediterranean diet , dietary approaches to stop hypertension , Alternate Healthy Eating Index , overall plant-based diet index , and adherence to national dietary guidelines . However, poorer diet quality increased the risk of gestational diabetes mellitus, including a higher dietary inflammatory index  and overall low-carbohydrate diets . After meta-regression, subgroup, and sensitivity analyses, the results remained statistically significant.Before and during pregnancy, higher diet quality reduced the risk of developing gestational diabetes mellitus, whereas poorer diet quality increased this risk.https://www.crd.york.ac.uk/PROSPERO/, identifier: CRD42022372488. Gestational diabetes mellitus (GDM) is the most prevalent medical illness in pregnancy and is defined as glucose intolerance of varying degrees, with onset or first detection during pregnancy . The aveGDM has many influencing factors, including race or ethnicity, family history of diabetes mellitus, age at delivery, obesity, overweight, and lack of exercise ; dietaryhttps://www.crd.york.ac.uk/PROSPERO/).The study protocol was registered in PROSPERO  Population: women before or during pregnancy who were involved in studies related to diet quality and GDM; (2) Exposure: studies that included diet quality as the exposure of interest, such as the Mediterranean diet (MD), dietary approaches to stop hypertension (DASH), Alternate Healthy Eating Index (AHEI), or other diet quality indices or scores; diet quality indices or scores referenced were based on established national or regional dietary guidelines; (3) Outcome: GDM; (4) Study design: cohort and case-control study; (5) Other inclusion criteria: results reported as the odds ratio (OR) or risk ratio (RR) with a 95% confidence interval (CI); studies using multiple dietary assessment methods were included.The exclusion criteria were as follows: (1) studies with no diet quality scores but only described dietary patterns, such as clusters, factors or reduced rank regression analysis; (2) studies that examined single nutrients, foods, or food groups; (3) randomized controlled trials, cross-sectional and qualitative studies; (4) studies involving animals; (5) unpublished data and gray literature, including conference abstracts, papers, and patents.http://www.prisma-statement.org/PRISMAStatement/FlowDiagram.aspx) were created to detail the inclusion/exclusion process.After duplicate removal, titles and abstracts were screened and full-text articles were obtained for further assessment. Study selection was independently conducted by two reviewers (Gao and Zheng). Any disagreement between the reviewers was discussed with a third reviewer (Jiang), who specialized in studying women's diets during the perinatal period. The PRISMA flowchart  . The NOSThe overall quality of evidence for the prevalence of GDM in the diet quality of included studies was assessed using GRADEpro Guideline Development Tool (GDT) software based on the principles of Grading of Recommendations, Assessment, Development, and Evaluations (GRADE) . To asseData extracted from each study included the first author, country, study design, follow-up duration/time, sample size, participants, dietary assessment tools, diagnostic criteria, key findings, OR/RR, and adjustment variables. For studies providing multiple estimates, we used the most complex model . If there was any disagreement during the data extraction, two researchers (Gao and Zheng) reviewed the full text and discussed it with a third reviewer (Jiang).pc is the usual occurrence rate without treatment   :Q-test and I2 index. When I2 ≥ 50, the random effects model was used; otherwise, the fixed effects model was used.ORs were log-transformed  for analysis. Between-study heterogeneity was examined using the P < 0.05, subgroup analysis was used for further exploration.For the purposes of the study, we conducted a subgroup analysis of the dietary assessment tools, types of participants (pregnancy or pre-pregnancy), and study design (cohort or case-control study). If significant heterogeneity remained after the analyses, we used subgroup analysis on study quality (low or moderate risk of bias) and country (developed or developing country) to identify the source of heterogeneity. Meta-regression analysis was used to identify the impact of adjustment variables on the study results (if there were more than 10 included studies); if Publication bias was assessed using Begg's and Egger's tests if more than 10 studies used the same diet quality assessment tool. Sensitivity analyses were performed to confirm the stability of the overall results . All staAfter removing duplicates, we screened the titles and abstracts of 2,429 articles for relevance. A total of 144 studies were identified as potentially eligible, and 19 were ultimately included in this systematic review. No additional articles were identified in the reference list. A flow chart is shown in n = 3) (n = 8) (n = 5) (n = 3)  , 42, 45  (n = 8) , 46, 48  (n = 5) , 43, 47, (n = 3) , 31, 33.n = 14). A total of eight diet quality assessment tools were included: one for Group A  and the other for Group B . We systematically evaluated these two groups separately. Group A included the MD, DASH, AHEI, overall plant-based diet index , and dietary guidelines (including China and Iceland); Group B included overall dietary inflammatory index  and overall low-carbohydrate diets . Three studies simultaneously used three diet assessment tools , and only a few studies adjusted for gestational weight gain (GWG), alcohol use, and socioeconomic status . We upgraded one level in the influence of residual plausible confounding factors because of the use of the most complex model. No serious issues were observed in the risk of bias, indirectness, and imprecision domains. The publication bias domain was downgraded, and there were no other additional factors. The overall GRADE recommendation in Group B (seven cohort) was “low-quality” which indicates that the true effect may be substantially different from the estimate of the effect; however, the rest were “very low-quality” which indicated that “any estimate of effect observed is very uncertain” , DASH , AHEI , MD , and dietary guidelines  were all inversely associated with the risk of GDM. The results of subgroup analysis by participants showed that high-quality diet was inversely associated with GDM in both pregnancy and pre-pregnancy . A subgroup analysis based on the study design indicated that heterogeneity could not be eliminated (I2 = 71%). A subgroup analysis of countries and study quality was conducted to determine the main parameters involved in heterogeneity. After stratification by country, between-study heterogeneity was removed in both subgroups (I2 = 29.9%); however, heterogeneity could not be eliminated through stratification of study quality . The meta-regression analysis showed that the adjustment variables had an impact on the study results . SubgrouMD and DASH were the most used diet quality assessment tools in the included studies (< 5 studies), and the significance of funnel plot asymmetry could not be tested. The results of the sensitivity analysis performed on Group A showed that the results of the systematic evaluation were reliable .I2 = 0.0%, the fixed effects model; I2 = 0.0%; OR: 1.37, 95% CI: 1.21–1.57, I2 = 24.5%, respectively). Subgroup analyses were conducted for pregnancy and pre-pregnancy, and the results indicated that both were positively associated with GDM . Subgroup analysis based on study design indicated that both cohort and case-control were positively associated with GDM  indicated that there was a significantly positive association between poor diet quality and GDM  with a total of eight diets, demonstrated that higher quality diet  before or during pregnancy reduced the risk of GDM, whereas poorer diet quality (higher DII or LCD diet) was associated with a high risk of GDM.Although the exact molecular mechanism remains to be elucidated, the results of this study are biologically plausible. As shown in Studies have shown that dietary patterns  are associated with a lower risk of GDM before and/or during pregnancy . HealthiIn contrast to dietary patterns, dietary quality assessment can combine large amounts of dietary data into a practical dietary indicator, thereby increasing the feasibility of translating food intake into daily food consumption and providing visualization of the intake of different food groups . StudiesDonazar-Ezcurra et al.  found thSome limitations of this meta-analysis should be considered. First, studies were observational, making causal inferences difficult. Although the studies adjusted for some confounders, the possibility of residual confounding cannot be ruled out completely. Most studies accounted for maternal age, BMI, and family history of diabetes; however, most of the included studies did not adjust for GWG, socioeconomic or alcohol status, previous macrosomia, or polycystic ovary syndrome, which may be important risk factors for GDM  Supplem. SecondlWe suggest the following recommendations for future studies. During the sampling and survey phases, studies should be conducted in diverse populations with varying racial/ethnic backgrounds, socioeconomic status, BMI ranges, and diet culture. Research should improve the collection methods of food intake, which can combine contemporary Internet technology such as applets and real-time recording, to capture the complexity of dietary habits more accurately. During the design and analysis phase, appropriate analytical methods should be used, and adjustments for covariates should be demonstrated through causal considerations and graphs, particularly for physical activity, gravidity, and socioeconomic status. Comparability studies should be improved by increasing the uniformity of the timing of dietary assessments and the outcomes measured. When possible, adequately powered randomized controlled trials should be conducted to support better causal inferences. In addition, one should investigate what are the micronutrient levels in women of childbearing age or in pregnant women with high or poor dietary quality.In conclusion, positive association between better maternal diet quality during pregnancy and functioning of the child. However, only a few studies have examined the relationship between diet quality and GDM. This study found that higher diet quality  before or during pregnancy reduced the prevalence of GDM; while poorer diet quality (higher DII or LCD) increased the risk of developing GDM. And then, the assessment of diet quality can have the potential to be a quick and easy way to screen for dietary habits associated with GDM before or during early pregnancy. Further studies are necessary to ascertain the relationship between food quality and GDM.The original contributions presented in the study are included in the article/XG, QZ, and XJ conceived and designed the experiments. XG and QZ performed the experiments and wrote the paper. XG, YP, and YL analyzed the data. XC contributed materials and analysis tools. All authors read and approved the final manuscript prior to submission.


Background: Reduced DNA repair capacity in nucleotide excision repair (NER) pathways owing to genetic variant may influence cancer susceptibility. According to published studies, variants of NER genes associations with colorectal cancer (CRC) risk were inconclusive. Thus, this meta-analysis aimed to explore the possible association. A trial sequence analysis (TSA) analysis was performed to control the risk of false positive or false negative.Methods: PubMed, Web of Science, Embase, Cochrane Library, China National Knowledge Network (CNKI), Wanfang Database and Scientific and Technical Journal Database (VIP) were searched to identify relative studies until April 2022. The association was assessed by odds ratio (OR) in Allele, homozygous, heterozygous, dominant, recessive, and over-dominant models. In addition, Begg’s and Egger’s tests, sensitivity analysis, subgroup analysis and TSA analysis were performed.Results: A total of 29 studies were eventually included in the meta-analysis, including 12,153 CRC patients and 14,168 controls. It showed that excision and repair cross complementary group 1 (ERCC1) rs11615 CC genotype decreased the risk of CRC, compared with TT genotype . For ERCC1 rs3212986, the significant impact was detected on increased the risk of CRC in the allele , homozygous , dominant  and recessive  models, especially in the Asian population. The results revealed the association of ERCC2 rs1799793 A allele with a higher risk of CRC . It also showed that ERCC5 rs17655 increased CRC risk in the allele , homozygous , heterozygous , dominant  and over-dominant  models, especially among Asians.Conclusion: This meta-analysis based on current evidence suggests that the significant association was observed between ERCC1 rs11615, ERCC1 rs3212986, ERCC2 rs1799793, and ERCC5 rs17655 and CRC susceptibility. However, given the limited sample size and the influence of genetic background, studies of a larger scale and well-designed are required to confirm the results. Globally, colorectal cancer (CRC) has the third highest incidence and second highest mortality rate, with a 5-year relative survival rate of 65% . By 2030XPC), XPA, excision and repair cross complementary group 1 (ERCC1), ERCC2/XPD, ERCC4/XPF, ERCC5/XPG  Statement .The terms used in search strategy were “colorectal OR colonic OR rectal OR colon OR rectum,” “cancer OR carcinoma OR neoplasms OR tumor,” “Single Nucleotide Polymorphism OR Polymorphism OR SNP OR variant OR variation,” and “nucleotide excision repair OR NER OR DNA repair OR excision repair.” In addition, studies published as of 6 April 2022 were extracted from PubMed, Cochrane Library, Web of Science, Embase, China National Knowledge Network (CNKI), Wanfang Database, and Scientific and Technical Journal Database (VIP) to meet the search strategy, regardless of language restrictions. Detailed strategies are shown in Inclusion criteria for the study were as follows: 1) It was a case-control study; 2) The difference in SNP frequency between CRC patients and healthy controls was compared; 3) The studied SNPs were NER gene SNPs; 4) Cancer risk is the outcome; 5) Cases were serum samples from CRC patients who did not receive chemotherapy; 6) The control group included healthy individuals with different age, sex, country, and tumor stage and patients with non-malignant disease.Two authors (Chuncheng Yi and Tiandong Li) independently extracted the literature. Data extracted from individual papers included: author, publication year, country, ethnicity, sample size, control type, gender composition, age at diagnosis, and details of target SNPs, including genotyping method, genotype frequency, and source from Hardy-Weinberg equilibrium (HWE).The quality of the studies that met the inclusion criteria was assessed. Two reviewers assessed the methodological quality of each study against the Newcastle-Ottawa scale (NOS) for case-control study . Scores 2 tests. I2 < 25% was considered to be of low heterogeneity, 25%–50% to be of moderate heterogeneity and 50% was considered a high level of heterogeneity. p < 0.05 or I2 > 50% was considered significant for heterogeneity, and a Dersimonian-Laird random effect model was applied if heterogeneity was not detected; otherwise, a Mantel-Haenszel fixed effect model was applied. Subgroup analyses were performed based on ethnicity and source of control to explore potential heterogeneity. A sensitivity analysis was performed to assess the stability of the results by excluding each study individually. The funnel plot evaluated whether publication bias was present (both Begg’s and Egger’s tests). For all statistical analyses, Stata version 14.0 software was used .A 95% confidence interval (95% CI) and odds ratio (OR) were utilized to determine the association between SNPs in NER gene and risk of CRC, and OR was analyzed using the Z test. Six different genetic models were used to test the association, including allelic, homozygous, heterozygous, dominant, recessive, and over-dominant models. Heterogeneity was assessed by Q test and Ihttp://www.ctu.dk/tsa).To control the risk of false positive or false negative, a trial sequential analysis (TSA) method was performed. If the curve exceeds the traditional boundary but does not cross the TSA boundary, the prompt may make a false positive error; If the curve intersects TSA boundary, the meta-analysis results are stable, even if required information size (RIS), which is the target value needed to make a reliable statistical inference, is not reached; If the curve does not intersect with both the traditional and TSA bounds, no positive or negative conclusion can be drawn; If the curve intersects an invalid line, the hint is meaningless. We performed TSA with a type I error risk of 5%, a power of 90% and a relative risk reduction of 15%, using TSA version 0.9.5.10 software . The result of quality assessment is shown in The flow chart of this study is shown in Upon analysis of these eight NER-SNPs, we found that four of them were significantly associated with CRC risk. Detailed results are shown in ERCC1 rs11615 homozygous CC genotype, but not the heterozygous CT genotype, decreased the risk of CRC, compared with the wild-type TT genotype    , however= 0.002) . The str= 0.002) .ERCC2 rs1799793, our meta-analysis showed that compared with G allele, A allele was associated with an increased risk of CRC   (For = 0.023) . And the= 0.023) . About t= 0.018) .ERCC5 rs17655 and CRC risk were noted under five genetic models. The correlation results under allele model was OR = 1.104, homozygous model was OR = 1.164, heterozygous model was OR = 1.271, dominant model was OR = 1.241 and over-dominant model was OR = 0.828 (The results demonstrated that significant associations between < 0.001) . Moreove< 0.001) .ERCC4 rs1800067 variant and CRC susceptibility in the six genetic models, nor in the stratified analysis. For XPC rs2228001 and rs2228000 variants, although no significant associations exist in the overall meta-analysis, ethnic subgroup analysis showed that significant correlations were noted in reducing or increasing the risk of CRC in Asian, Caucasian and African populations under different models. The detailed results are shown in No significant association was observed between ERCC1 rs3212986 had high heterogeneity . We then assessed the source of heterogeneity by ethnicity and control source. The results showed that heterogeneity among the Asian and population-based control subgroups decreased. Then, sensitivity analyses were performed to assess the stability of the results by deleting each study in sequence. The results showed that none of the studies significantly changed the combined OR, indicating that the combined OR of this variant was robust. Heterogeneity was low for the remaining meaningful SNPs, and the sensitivity analysis results were robust. Under the allelic model, Begg’s funnel plot and Egger’s regression test were performed to estimate publication bias. The results showed no significant asymmetry of funnel plot shape , and BegERCC1 rs11615 variant and CRC risk under homozygous model, similar to ERCC2 rs1799793. For ERCC1 rs3212986 variant in the four genetic models, the Z curve only crossed the traditional boundary, and the sample size did not reach RIS, indicating that there was a high possibility of false positives in the results. Therefore, more subsequent studies are needed for further verification to make this association valid . In future research, the following aspects should be paid attention to. Firstly, there may be a need for further research on different SNP topics in the NER gene, involving broader population regions and sample sizes. Secondly, gene-environment interactions and gene-gene interactions should also be considered in subsequent studies. Finally, functional studies are required to elucidate the underlying mechanisms of NER genetic variants in tumorigenesis.In conclusion, our meta-analysis of studies in different populations indicated that


The risk difference was quantified to be 0.037, 0.030, 0.017, and 0.009, respectively. Conclusions: We quantified the relative risk of numerous adverse events associated with the use of ACE inhibitors in a variety of demographics. This information can help healthcare providers be fully informed about any potential adverse consequences and make appropriate suggestions for their patients requiring ACE inhibitor therapy.Background: Although angiotensin-converting enzyme (ACE) inhibitors are among the most-prescribed medications in the world, the extent to which they increase the risk of adverse effects remains uncertain. This study aimed to systematically determine the adverse effects of ACE inhibitors versus placebo across a wide range of therapeutic settings. Methods: Systematic searches were conducted on PubMed, Web of Science, and Cochrane Library databases. Randomized controlled trials (RCTs) comparing an ACE inhibitor to a placebo were retrieved. The relative risk (RR) and its 95% confidence interval (95% CI) were utilized as a summary effect measure. A random-effects model was used to calculate pooled-effect estimates. Results: A total of 378 RCTs fulfilled the eligibility criteria, with 257 RCTs included in the meta-analysis. Compared with a placebo, ACE inhibitors were associated with an significantly increased risk of dry cough (RR = 2.66, 95% CI = 2.20 to 3.20,  Angiotensin-converting enzyme (ACE) inhibitors are among the most widely prescribed drugs in the world, with hundreds of thousands of patients exposed to this drug class every year ,2. Ever ACE inhibitors can cause adverse drug reactions, such as dry cough ,12, and In this study, we followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) 2020 guidelines  and the An initial literature search was performed in three medical databases, namely PubMed, Web of Science, and Cochrane Library (last search: 15 November 2020), supplemented by a review of the reference lists of relevant articles for potentially eligible studies. Search strategies were conducted with the terms related to ACE inhibitors  AND placebo, with no language restriction. All articles that were deemed relevant based on the title and abstract screening were retrieved and assessed with respect to eligibility criteria: (1) a study involving human subjects, (2) a study investigating the effect of an ACE inhibitor, (3) a study having a placebo-control group, and (4) a study having a report of adverse outcomes. Included articles were limited to an RCT  design. Observational studies, review articles, nonhuman studies , and other types of articles  were excluded. Studies were independently selected by at least two review authors, and the decision to include or exclude was arbitrarily based on consensus.We applied a hybrid approach consisting of prespecified adverse outcomes of interest (confirmatory approach) and additional common adverse outcomes being later identified (exploratory approach). Prespecified adverse outcomes of interest included four well-known adverse outcomes of ACE inhibitors, i.e., dry cough, hypotension, hyperkalemia, and angioedema. Additional adverse outcomes (either self-reported or actively sought) that were identified in more than three studies while conducting the systematic review were also included in the meta-analysis. The review authors did not group adverse outcome data in a composite measure. The data were not extracted as “0” unless reported as such in the original publication. A general statement indicating the absence of the event  was not extracted as a “0” event and such a statement was not included in further analysis. In the case that there were multiple reports from the same trial, the most complete and relevant dataset was used for analysis. For any unclear data, we made several attempts to e-mail the corresponding authors to ask for clarification.Five domains of bias  were assessed following the most updated tool for assessing the risk of bias in RCTs (RoB2) . Each RCQ test and the I2 statistic; I2 > 75% suggests high heterogeneity, whereas I2 < 40% indicates that statistical heterogeneity might not be important [I2 > 50%) was explored using subgroup analyses and/or meta-regression, as appropriate, based on prespecified factors, i.e., drug name, drug dosage, and treatment duration. Sensitivity analyses were performed, with respect to the risk of bias, by excluding studies with a high risk of bias.For each adverse outcome, the extracted data were tabulated in a 2 × 2 contingency table. The relative risk (RR) and its corresponding 95% confidence interval (95% CI) were used as a summary effect measure. Pooled-effect estimates were calculated by means of a random-effects model, which reflected the systematic variation in estimates among studies. A funnel plot was applied to explore the potential of small-study effects, including publication bias, followed by the application of a linear regression test when there were 10 or more RCTs in a meta-analysis . Statistmportant ,29. Subsp value of < 0.05 was regarded to indicate statistical significance. A package ‘meta’ version 4.20-1 was executed to analyze meta-analyses. The package ‘ggplot2’ version 3.3.5 was used to construct heatmap visualization, and package ‘circlize’ version 0.4.13 was used to generate a chord diagram.All statistical tests were performed in RStudio version 1.3.1093  for meta-analyses and data visualization, and a n = 41,263, from 65 studies), perindopril , ramipril , and enalapril  , followed by cerebrovascular disease  and endocrine disorders   . The effstudies) . The medstudies) . Qualitap < 0.001), hypotension , and hyperkalemia . However, we found no statistically significant association between ACE inhibitor exposure and the incidence of angioedema . Among the prespecified adverse outcomes of interest, the heterogeneity was low for all the outcomes, except for dry cough (I2 = 81.5%)  (p = 0.005) (p < 0.001), 0.030 , and 0.009  for dry cough, hypotension, and hyperkalemia, respectively.Compared with a placebo, ACE inhibitors were significantly associated with an increased risk of dry cough  . Substan= 81.5%) . Meta-rectively) . The fun= 0.005) . Sensiti= 0.005) . The risp < 0.001). The heterogeneity was low (p = 0.106). No significant changes in the summary effect estimates were found when sensitivity analyses were applied (p < 0.001). Overall, drug discontinuation probably due to adverse drug reactions did not statistically significantly differ between the ACE inhibitor group and the placebo group . Meta-regression did not find a dose-dependent relationship between ACE-inhibitor-dose levels and drug discontinuation . Additionally, drug discontinuation was not significantly associated with treatment duration .An exploratory approach identified one more adverse outcome associated with ACE inhibitor exposure, i.e., dizziness , that is, with approximately 28 patients (95% CI = 23–34) treated with ACE inhibitors, 1 patient would present with dry cough attributable to an ACE inhibitor. Based on meta-regression and subgroup analysis, ACE-inhibitor-related dry cough was not dose- or duration-dependent; perindopril and ramipril yielded the highest significant risk of dry cough , while enalapril yielded the lowest risk of dry cough (RR = 1.23). The mechanism of ACE-inhibitor-induced dry cough is perceived to be related to a bradykinin-induced sensitization of airway sensory nerves and an accumulation of substance P in the respiratory tracts .We observed increased risks of hypotension (RR = 1.98) and hyperkalemia (RR = 1.24) among individuals taking ACE inhibitors. The absolute risk increase was 0.030 and 0.009, respectively; that is, with approximately 33 and 111 patients who are treated with ACE inhibitors, 1 individual patient would present with hypotension and hyperkalemia, respectively, attributable to ACE inhibitor usage. These adverse drug reactions are attributable to the mechanism of action of ACE inhibitors related to reduced angiotensin II formation ,33. It wIn our meta-analysis, ACE inhibitors were found to be statistically significantly associated with dizziness. Of every 59 patients (95% CI = 41–109) who are treated with ACE inhibitors, there will be approximately 1 individual complaining of dizziness attributable to ACE inhibitor treatment. Our finding is consistent with a reverse association study in which the authors reported ACE inhibitors to be one of the most common drug classes that were associated with dizziness among patients who presented with dizziness with an unknown cause . Althougp = 0.160). The discrepancy between our findings and previous evidence may be owing to angioedema occurring more commonly among certain populations  [In contrast with previous evidence ,39, our atients) ,41,42, watients) ,44. As datients) ,46,47, satients) .There is a possibility that our findings based solely on published RCTs may misestimate the risk estimates of ACE-inhibitor-related adverse outcomes ,50. The In conclusion, in the present systematic review and meta-analysis, we characterized and quantifies the relative risk of several adverse outcomes due to ACE inhibitors across a diverse range of populations. Treatment with ACE inhibitors puts individuals at increased risk of dry cough, hypotension, dizziness, and hyperkalemia by 2.66, 1.98, 1.46, and 1.24 times, respectively. The overall findings of the present study help in guiding informed decisions about the management of diseases requiring ACE inhibitor therapy. These findings can also be used to advise patients requiring ACE inhibitor therapy.