Diabetes medications may modify the risk of certain cancers. We systematically searched MEDLINE, Embase, Web of Science, and Cochrane CENTRAL from 2011 to March 2021 for studies evaluating associations between diabetes medications and the risk of breast, lung, colorectal, prostate, liver, and pancreatic cancers. A total of 92 studies  were identified in the systematic review, involving 171 million participants. Inverse relationships with colorectal  and liver cancers  were observed in biguanide users. Thiazolidinediones were associated with lower risks of breast , lung  and liver  cancers. Insulins were negatively associated with breast  and prostate cancer risks . Positive associations were found between insulin secretagogues and pancreatic cancer , and between insulins and liver  and pancreatic cancers . Overall, biguanide and thiazolidinedione use carried no risk, or potentially lower risk of some cancers, while insulin secretagogue and insulin use were associated with increased pancreatic cancer risk. Epidemiological evidence suggests that diabetes is associated with a higher risk for many cancers, from 20 to 30% elevated breast or colorectal cancer to 97% elevated risk of intrahepatic cholangiocarcinoma or endometrial cancer3. These associations likely reflect shared etiology. For instance, tobacco use, physical inactivity, overweight, and poor diet are closely linked to cancer risk and progression4 and risk of diabetes7. Several biological mechanisms have been proposed to explain why diabetes may be associated with cancer, including inflammation, hyperglycemia, and hyperinsulinemia9. Furthermore, medications for diabetes may modify the association. Multiple epidemiologic studies have evaluated the association of diabetes medications with the risk of site-specific cancers in patients with diabetes11.The prevalence of type II diabetes, herein referred to as diabetes, continues to grow and, globally, will affect 7.7% (439 million) of adults aged 20 to 79 years old by 203014 or the effects of multiple medication classes on overall cancer15. No systematic reviews have synthesized multiple associations among different diabetes medication classes and cancer sites in patients with diabetes, which limits our capacity to analyze evidence in the context of various mechanisms involved with diabetes treatment and underlying causes of cancers. To this end, we conducted a systematic review and meta-analyses of the latest randomized controlled trials (RCTs) and observational studies (cohort and case–control studies) published in the last 10 years that evaluated the effect of diabetes medications, namely biguanides, incretin-based medicines, alpha-glucosidase inhibitors (AGIs), insulin secretagogues, thiazolidinediones, and insulins, on the risk of developing any of study cancers, specifically breast, lung, colorectal, prostate, liver, or pancreatic cancers.Previous systematic reviews either synthesized the effect of a single medication class on multiple cancer sitesThe literature search and study selection process are summarized in Fig. All three RCTs were categorized as having a high risk of bias due to either lack of allocation concealment or potential selective reporting Table . Among tOverall, colorectal (n = 57) and pancreatic cancers (n = 42) were the most commonly evaluated outcomes across included studies Table . A cohorOf the studies that reported specific diabetes medication classes, the majority studied biguanides and insulins, while only six studies evaluated the effect of AGIs on cancer risk. Most studies (67%) were of a cohort design. The proportions of studies with a low risk of bias were similar across diabetes medication classes. The majority of studies of biguanides and insulin secretagogues involved populations from Western regions, while the majority of studies of AGIs were conducted in populations from Asian regions.Individual associations among medication classes and cancer sites and subgroup analysis results are summarized in Tables Effects were largely consistent across studies with different designs, quality levels, and population regions, though there are some notable differences. Subgroup analyses revealed that the positive association between insulin use and liver cancer was restricted to studies conducted in populations from Western regions  across the included studies.16. The potential protective association of biguanides with liver cancer might be further attributed to decreased lipogenic enzyme gene expression and suppressed lipogenesis17. By stimulating endogenous insulin production, insulin secretagogues increase insulin-like growth factor-1 levels which may promote pancreatic cancer development by interfering with cell metabolism and stimulating cell proliferation18. Thiazolidinediones have been shown to suppress breast cancer cell proliferation, stimulate apoptosis, and impede tumor angiogenesis with PPAR-γ ligands19. Thiazolidinediones-induced PPAR-γ activation has also been shown to hinder lung tumor progression through G0/G1 cell cycle arrest20. The inverse association observed between thiazolidinediones and liver cancer may be attributed to p27Kip1 protein accumulation which is associated with reduced liver cell growth21.Several biological mechanisms have been speculated to contribute to cancer modifying effects of diabetes medications. The protective effects of biguanides on colorectal cancer may be attributed to their ability to regulate upstream and downstream molecular mechanisms involved in cellular metabolism and energy homeostasis, cell cycle arrest, oxidative stress, inflammation, and apoptosis22. The cancer-promoting effect of insulins on the liver might be partly explained by portal circulation, which exposes the liver to high levels of insulin in people who have insulin-resistant diabetes23. In addition, insulin use has growth-promoting and mitogenic effects on pancreatic cancer cells, which may lead to pancreatic cancer development or acceleration24. However, the higher risk of pancreatic cancer might also be attributable to severe diabetes conditions rather than insulin use.A protective effect of insulin was observed on breast cancer in the review, but the etiology is unclear. The lack of adjustment for mammography screening suggested detection bias might be a concern. Our finding of lower prostate cancer risk in men with diabetes using insulins is similar to a previous study. Men who were on insulins were more likely to have severe diabetes which might be linked to lower prostate-specific antigen levels and reduced risk of prostate cancer25 found that individuals with diabetes had an overall increased cancer risk, our study highlights the impact of specific diabetes medications on site-specific cancer risk. In alignment with the systematic review and meta-analysis by Noto et al. 26, our findings underscore the potential protective effect of biguanides, such as metformin, in reducing the risk of colorectal and liver cancers. Similarly, the protective effect of metformin in our study echoes DeCensi et al. 27, implying a robust consensus across studies regarding this association. Our findings that insulin and insulin secretagogue use can be linked to an elevated risk of certain cancers align with the results from Karlstad et al. 28 and Singh et al. 29. These consistent findings across studies highlight the potentially harmful effects of these medication classes on cancer risk, meriting careful clinical consideration. In addition to confirming results from previous systematic reviews, our review broadens the understanding by incorporating more recent studies, diverse geographical regions, and differing study designs, thus enhancing the comprehensiveness and applicability of the findings.Although Noto et al. Our meta-analysis reveals associations between diabetes medication use and cancer risk that were generally consistent across diverse study designs, quality levels, and population regions. However, regional differences were discovered, such as the stronger association between insulin use and liver cancer risk in populations from the Western regions. The inconsistent association between insulin secretagogues and colorectal cancer and biguanides and pancreatic cancer, which varied by study design, underscores the need to interpret results cautiously. Our findings indicate that geographical factors and study design might influence observed associations, necessitating additional research to enhance our understanding.30. Our approach, considering different classes of medications separately across individual cancer sites, is an important advance over previous literature reviews15 that have grouped all diabetes medications and/or cancer sites together. The large sample size of included studies augmented the precision of summary effect estimates, and the inclusion of large numbers of individuals from Asian and Western regions enhances the generalizability of our findings.This review represents the latest examination of associations among all currently prescribed classes of diabetes medications and the risk of cancer across multiple sites. A myriad of new diabetes medications has become available over the past two decadesOur review has some limitations. Evidence of heterogeneity in summary risk estimates was observed for biguanides and insulins across all cancer sites, but subgroup analyses did not reveal clear sources of this heterogeneity. It is possible that heterogeneity reflects differences in study population demographics such as varying proportions of males/females and differences in age or unmeasured confounding within studies, for example, few studies adjusted for lifestyle factors like nutritional status that may confound associations. It also appeared that study design impacted results. For instance, the use of insulin secretagogues showed a more prominent effect on liver and pancreatic cancers in case–control studies than in cohort studies. When the majority of studies were case–control studies, differences observed in results between cohort and case–control studies might be attributed to recall bias. Effects of sodium-glucose cotransporter 2 inhibitors on cancer risk and AGIs on breast and prostate cancer could not be explored due to the absence of data. In general, few studies of cancer risk in association with AGIs and incretin-based medicines were identified. Differences in the standard of care for diabetes across jurisdictions and the use of combination drug therapies may have impacted results. Because the severity of diabetes is typically a driver of specific medication use, the possibility of confounding by indication cannot be ruled out in the individual studies that were included in this review. We did not evaluate dose–response relationships by incorporating the duration of diabetes medication use because it is beyond the scope of this review.In conclusion, various diabetes medications might be linked to the risk of study cancers; however, the strengths of associations showed variation to some extent across studies with different designs, quality levels, and population regions. On the other hand, the preventive effect of biguanides against colorectal and liver cancers was consistent across different studies. A potential cancer-promoting effect of insulins on the liver was only found in populations from Western regions, whereas the increased risk associated with insulin on the pancreas was only found in populations from Asian regions. Although further studies are required, these findings suggest that it may be important to weigh the potential harms of insulin among patients with diabetes who are at high risk of liver or pancreatic cancers due to family history or other risk factors. The choice of medication for patients with diabetes may incorporate individuals’ susceptibility to individual cancers in conjunction with established clinical considerations. Future studies may further consider the duration of medication use and dose–response relationships to consolidate current knowledge in observed associations with study cancers.31 statement. The completed PRISMA checklist can be found in Supplementary file The search adhered to the guidelines provided by the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA)32, and thus it is critical to understand possible risk factors. As a result of the high prevalence, we also expected to identify a larger body of evidence for breast, lung, colorectal and prostate cancer from which to synthesize findings. Although cancers of the liver and pancreas are less common, they have high fatality rates. Moreover, liver and pancreatic cancers were included in this review due to biological plausibility of diabetes medication and liver and pancreatic cancer, reflecting the roles of the liver and pancreas in blood glucose regulation and diabetes. Studies using a cross-sectional design were excluded as they could not provide clear evidence of temporality. In this study, we reviewed the results of RCTs, cohort, and case–control studies. Reviews, commentaries, opinion pieces, letters, and case reports were also excluded.We included original human experimental and observational studies that evaluated the associations between diabetes medications and six site-specific cancers: breast, lung, colorectal, prostate, liver, and pancreatic cancers. The first four cancers are the most prevalent cancers worldwide The study population for breast cancer was restricted to female biological sex due to the rareness of breast cancer in males. People with prevalent cancer or a previous history of study cancers were excluded. Studies focusing on pediatric or adolescent populations were also excluded. No other restrictions were placed on the study population.In RCTs, diabetes medications were randomly assigned. For observational studies, the determination of the exposure to diabetes medications was made on the basis of medical records, pharmacy prescription records, or insurance claims databases. Self-reported exposure was also included, recognizing its inherent limitations. Studies were excluded if they did not provide explicit methodologies for establishing exposure to diabetes medication.Studies without an appropriate comparator group were excluded. In addition to people receiving no treatment or placebo, for each diabetes medication of interest, people using one or more other diabetes medications were considered to be eligible as controls. Control groups were from population-based and hospital settings in observational studies and in RCTs, respectively.Cancer incidence or mortality was determined through a national record system, cancer registry or death certificates. Eligible measures of the effect included odds ratio (OR), relative risk (RR), and hazard ratio (HR). The absence of these specific effect measures led to exclusion.This systematic review focuses on research published in the last ten years. Searches were limited to publications dating from January 2011 onwards. The timeframe was chosen to reflect advances in pharmaceutical treatment of diabetes; the introduction of new medications e.g., GLP-1 receptor agonists and SGT-2 inhibitors with distinct mechanisms of action from existing diabetes medication classes. The focus on recent literature is thus more likely to capture evidence on diabetes medications currently prescribed and have greater clinical relevance. No language restrictions were imposed. After the titles/abstracts screening, the full text of selected studies in languages other than English was translated using Google Translate.With the assistance of an institutional librarian, we systematically searched MEDLINE (Ovid), Embase (Ovid), Web of Science Core Collection, and Cochrane CENTRAL (Ovid). Search terms included controlled vocabulary and keywords related to various diabetes medications and cancer at multiple sites. Details of the search strategy for each database are provided in Supplementary file https://www.covidence.org). The study selection comprised two stages. Firstly, all titles and abstracts were screened by reviewers  for potentially eligible papers according to the above-mentioned inclusion and exclusion criteria. Studies that were excluded by more than one reviewer were excluded from this review. Secondly, full texts of all relevant studies were retrieved, and two reviewers  independently screened each study for eligibility. Reasons for exclusion were documented at the full-text screening stage.All titles and abstracts were imported to Covidence 34. If randomization, allocation concealment, and blinding were uncertain in an RCT, it was categorized as a high-risk study15. Three domains assessed in observational studies were the selection of study groups, the comparability of groups, and the ascertainment of exposure or outcome of interest. A score of less than 4 or more than 6 represented a high or low risk of bias, respectively15.The methodological quality of RCTs and observational studies was scrutinized independently by two reviewers and collated by the lead reviewer using Cochrane Collaboration’s tool for assessing the risk of bias35 to pool the data from two or more studies evaluating the same class of diabetes medication and the same site of cancer. Cochran’s Q was used to test the statistical heterogeneity of the included studies, with a significance level of 0.136. The magnitude of heterogeneity was measured using the I2 statistic37, with < 25% and > 75% corresponding to low and high heterogeneity, respectively. Analyses were performed to quantify individual associations among different medication classes and cancer sites. We performed subgroup analyses  to determine possible sources of between-study heterogeneity.We performed meta-analyses to obtain the pooled relative risk (RR) using Stata software version 17.0 . First, the findings were summarized in a narrative form, with textual description and tabulation to compare potential differences among studies. We employed a DerSimonian-Laird random-effects modelSupplementary Information 1.Supplementary Information 2.Supplementary Information 3.Supplementary Information 4.


Recently, four meta-analyses have explored the association between inflammatory bowel disease (IBD) and the risk of stroke. These studies have demonstrated that people with IBD may be at an increased risk of stroke. However, some limitations such as high heterogeneity and the lack of uniformity in the types of research, especially the reuse of some sample sizes, cannot be neglected. These factors reduce the credibility of their research conclusions. Therefore, we conducted a meta-analysis to explore this possible association.PubMed, Embase, and Web of Science were searched from inception to 30 June 2023. A random effects model with the generic inverse variance method was used in this meta-analysis. The Review Manager software was used to obtain all relative risks (RRs) and their 95% confidence intervals (CIs). Publication bias was tested, and sensitivity and subgroup analyses were conducted to explore possible heterogeneities.This meta-analysis included 12 cohort studies . Meta-analysis of these data has shown that IBD was associated with an increased risk of stroke . Our results were stable and robust in subgroup and sensitivity analyses.Our results suggest that IBD is associated with an increased risk of stroke. To reduce the incidence of stroke, patients with IBD are encouraged to undergo stroke risk assessments, especially for young female patients; assessing the risk of ischemic stroke is of particular importance. Prospective studies considering stroke subtypes, IBD severity and treatments, regions, and other confounding factors are needed to further explore the nature of each association.https://www.crd.york.ac.uk/PROSPERO/, identifier CRD42022373656. Inflammatory bowel disease (IBD) is a chronic inflammatory disease in the gastrointestinal tract, with Crohn's disease (CD) and ulcerative colitis (UC) being types of the condition. The epidemiology of IBD has changed significantly compared to the past. The highest rate of inflammatory bowel disease (IBD) is observed in the Western world, yet in newly industrialized countries in Asia, Africa, and Latin America, the incidence of IBD is increasing at a rapid rate –3. With According to the Global Burden of Disease 2019, stroke has become the second leading cause of death and the third leading cause of disability and death combined globally. From 1990 to 2019, the total number of stroke events rose by 70.0%, epidemic stroke increased by 85.0%, and stroke deaths increased by 43.0% , 7, resuStudies have revealed a close association between IBD and a range of neurological disorders, including Parkinson's disease, dementia, multiple sclerosis, and depression –19. GiveWe performed the systematic review and meta-analysis in accordance with the PRISMA   reportinWe conducted a comprehensive search of PubMed, Embase, and Web of Science from inception to 30 June 2023 to identify all studies related to the association between inflammatory bowel disease (IBD) and stroke. The following keywords were used in our search strategies: (inflammatory bowel disease OR ulcerative colitis OR Crohn's disease) AND . We conducted our search combining MeSH terms with text word searching, without any language restriction. The online The included studies met the following inclusion criteria: (1) cohort studies; (2) the exposure was IBD and the outcome was stroke appearing after the diagnosis of IBD; and (3) studies reported relative risk (RR), hazard ratio (HR), or incidence rate ratio (IRR) with 95% confidence interval (CI) for stroke incidence or provided the original data for the calculation. The exclusion criteria were as follows: (1) laboratory studies, abstracts, reviews, meta-analyses, comments, letters, and case reports; (2) articles without sufficient data; (3) IBD appearing after the diagnosis of stroke; and (4) the study did not provide sufficient data to calculate the RR, HR or IRR and 95% CI. Studies were independently screened by two reviewers who read titles and abstracts and obtained the full text of potentially relevant articles based on the inclusion and exclusion criteria. Any differences were resolved by consensus.We extracted the following information from each eligible study: the first author, the year of publication, districts, study subjects, sample size, follow-up duration, type of IBD and stroke, mean age or age of the group, sex, the proportion of males, and RR or HR (adjusted and unadjusted) with their 95%CI. Discrepancies were settled with mutual agreement by two reviewers who independently extracted the data from each included study.The Newcastle-Ottawa Scale (NOS) , which i2 statistic to evaluate heterogeneity among studies. Cochran's Q-test was used to evaluate the heterogeneity, P < 0.10 for the Q-test was considered statistically significant and I2 value was used to assess the degree of heterogeneity. If I2 values were <25%, we considered low heterogeneity, values in the range of 26%-50% were considered as having medium heterogeneity, and above 50% as having high heterogeneity. When the heterogeneity was >50%, it was necessary to employ a random effects model to combine the data; otherwise, a fixed effects model was used. Sensitivity/subgroup analyses were performed according to the type of IBD and stroke, race, age, and sex of study participants, follow-up period, and adjustment for confounders to explore the potential sources of heterogeneity. Additionally, the funnel plot and Egger's regression asymmetry were used to evaluate the potential of publication bias with the STATA/SE software .This study aimed to investigate the association between IBD and the occurrence of ischemic stroke, hemorrhagic stroke, and unclassified stroke. Meta-analyses were performed using the Review Manager 5.3 . RR and 95% CI from each study were calculated using a random-effects model, the generic inverse variance method of DerSimonian and Laird. We used Cochran's Q-test and IAs specified in the PRISMA flow diagram, there were 3,431 potentially relevant articles retrieved from the three electronic databases using the initial search strategy. A total of 1,269 duplicates were excluded. By reading the titles and abstracts, we obtained 68 records, of which 12 articles were finally included after reading the full texts .A total of 12 cohort studies –37 met on = 17, RR = 1.19, 95%CI:1.14–1.24, p < 0.00001 2 = 59%, p = 0.001, see n = 11, RR = 1.30, 95%CI:1.20–1.41, p < 0.00001; I2 = 65% n = 11, RR = 1.11, 95%CI:1.06–1.16, p < 0.00001; I2 = 36% n = 9, RR = 1.18, 95%CI:1.14–1.21, p < 0.00001; I2 = 0% n = 8, RR = 1.21, 95%CI:1.17–1.24, p < 0.00001; I2 = 73% n = 9, RR = 1.14, 95%CI:1.10–1.17, p < 0.00001; I2 = 0% n = 12, RR = 1.22, 95%CI:1.10–1.35, p = 0.0001; I2 = 80% n = 8, RR = 1.46, 95%CI:1.17–1.82, p = 0.0007; I2 = 63% n = 10, RR = 1.30, 95%CI:1.18–1.42, p < 0.00001; I2 = 68% n = 10, RR = 1.12, 95%CI:1.06–1.19, p < 0.0001; I2 = 31% n = 12, RR = 1.21, 95%CI:1.16–1.27, p < 0.00001; I2 = 64% n = 5, RR = 1.10, 95%CI:1.02–1.19, p = 0.02; I2 = 17% n = 15, aRR = 1.17, 95%CI:1.14–1.19, p <0.00001; I2 = 63% n = 4, cRR = 2.69, 95%CI:2.61–2.78, p < 0.00001; I2 = 99% n = 2, RR = 1.23, 95%CI:1.14–1.32, p < 0.00001; I2 = 22%; UC n = 2, RR = 1.14, 95%CI:1.03–1.27, p = 0.009; I2 = 76% n = 2, RR = 1.38, 95%CI:0.83–2.30, p = 0.21; I2 = 93%; UC n = 2, RR = 1.16, 95%CI:0.85–1.58, p = 0.35; I2 = 87% In all, 12 studies confirmed the link between IBD and stroke risk. A pooled analysis showed that IBD was significantly associated with an increased stroke risk (2 = 59%). Our results remained consistent after subgroup analyses by race, sex, age, and year of follow-up, underscoring the robustness of our results. Furthermore, our results are also in line with previous research findings  and the largest sample size , especially by including newer cohort studies , so we conducted subgroup analyses to identify the sources of this heterogeneity. However, after considering the IBD and stroke subtypes, age, follow-up time, sex, and race, we concluded that they were not the main sources of heterogeneity. In addition, to evaluate the stability of pooled results, we conducted a sensitivity analysis with the omission method ; the results of our meta-analysis for IBD and stroke were observed to be relatively stable by this method.Our results revealed considerable heterogeneity (random-effects RR = 1.19, 95%CI, 1.14–1.24; IThis study has some strengths and limitations that should be acknowledged. First, compared with previous meta-analyses on this topic, our research encompasses the newest and most extensive set of literature, and the estimated effect sizes were the same in all the included studies, which can minimize heterogeneity. Second, our team found that some articles , 38–41 fOur findings point to a link between IBD and an elevated risk of stroke. To lessen the risk of stroke, it is recommended that patients with IBD have a stroke risk evaluation, especially young female patients; assessing the risk of ischemic stroke is of particular importance. To gain a fuller understanding of the association between the two, further research is necessary that takes into account factors such as stroke subtype, IBD severity and treatment, region, and other potential confounders.The original contributions presented in the study are included in the article/J-SF and MW performed the literature search, drafted the manuscript, and designed the systematic review. MW and B-cS screened the literature. NC, Q-BZ, and J-SF extracted data and assessed the quality. M-JH and YL analyzed and interpreted the data. MW, J-SF, and NC were responsible for the research design, data analysis, and manuscript revision. All authors gave their approval for the submission of the final manuscript.


In this meta-analysis, findings from recent studies on the preventive properties of anthocyanins (ACN) against cardiovascular disease are summarized.MEDLINE, PubMed, Embase, Cochrane Library, and Google Scholar were searched and 2512 studies were found in a preliminary search. After screening of titles and abstracts, 47 studies met the inclusion criteria . Studies were excluded based on the following criteria: incomplete data; obscurely reported outcomes, or lack of control groups; and animal studies.The results showed that intervention with ACNs resulted in a significant decrease in body mass index ((MD),− 0.21; 95% CI, − 0.38, − 0.04; P < 0.001) and body fat mass . Pooled data comparing ACN with control showed statistically significant effect on fasting blood sugar and HbA1c. However, the reductions were significantly more in the subjects with type 2 diabetes and in those who used ACN as supplement/extract. The subgroup analysis test showed that there was a significant effect of ACN on triglyceride concentrations, total cholesterol, LDL-C and HDL-C concentrations in all subgroups of participants (with vs. without dyslipidemia at baseline) and intervention type (supplement/extract vs. food). However, we did not observe any significant effect on apo A and apo B concentrations.http://www.crd.york.ac.uk/Prospero (Registration no. CRD42021286466).ACN intake in the forms of natural foods and supplements can induce healthy changes in body fat mass, glycemic and lipidemic status and these effects are more prominent in the subjects with above-normal values. This meta-analysis was registered at The online version contains supplementary material available at 10.1186/s13098-023-01075-0. Cardiovascular diseases (CVD), comprising ischemic heart disease, stroke, heart failure, peripheral arterial disease, and a number of other cardiac and vascular conditions are still the leading contributors to disease burden worldwide. According to the World Health Organization (WHO) by 2025, over 28 million individuals will die from CVD annually , 2. TherOver the past decade, there has been an increasing interest in lifestyle and dietary interventions to lessen chronic disease risks. A huge body of evidence shows that fruits and vegetables intake is inversely associated with the risk of the most common chronic diseases including CVD . These eAnthocyanins (ACN) seem to be the major subgroup of flavonoids (pigments occurring in fruits and vegetables) having exclusive metabolic and biological behaviors , 7. TherA growing body of evidence indicates that consumption of ACN-rich products can reduce the risk of CVD, mainly in the subjects at risk , 11. Forhttp://www.crd.york.ac.uk/Prospero (Registration no. CRD42021286466).This systematic review and meta-analysis was planned, conducted, and reported according to the widely accepted quality standards  and was The interventions investigated in this review were giving ACN either as supplement or food in various doses and frequencies as compared with a control group. No limit was placed on the dose, type of vehicle or frequency at which supplements and/or fortified foods were taken.Outcome measures were cardiometabolic risk factors including:2), waist circumference (WC) (cm), body fat mass (FM)(%).Anthropometric measures: weight (kg), body mass index (BMI) (kg/mGlycemic indicators: fasting blood glucose (FBG) (mg/dL), fasting serum insulin (FSI) (µIU/mL), two-hour glucose tolerance test (mg/dL), homeostasis model assessment-estimated insulin resistance (HOMA-IR), glycated hemoglobin (HbA1c) (%).Lipid profile: serum triglyceride (TG) (mg/dL), total cholesterol (TC) (mg/dL), low density lipoprotein-cholesterol (LDL-C) (mg/dL), high density lipoprotein-cholesterol (HDL-C) (mg/dL), apolipoprotein A (Apo A) (mg/dL), apolipoprotein B (Apo B) (mg/dL).A comprehensive search of electronic databases was conducted for eligible trials up to September 2021. The following databases were covered: National Library of Medicine (PubMed), Scopus, Web of Science (WoS), and Cochrane Database of Systematic Reviews  using the following search terms in titles and abstracts: 1- (anthocyanin) AND , waist circumference (WC), fat mass (FM)) 2- (anthocyanin) AND (Glycemic indicators: fasting blood glucose (FBG), fasting serum insulin (FSI), two-hour glucose tolerance test, homeostasis model assessment-estimated insulin resistance (HOMA-IR), HbA1c), 3- (anthocyanin) AND (lipid profile: serum triglyceride (TG), total cholesterol (TC), low density lipoprotein-cholesterol (LDL-C), high density lipoprotein-cholesterol (HDL-C), apolipoprotein A (apo A), apolipoprotein B (apo B)).All of the studies were limited to English language and those in humans. All the references in the papers of the included studies were scanned for further relevant reports.all studies with a randomized clinical trial design;data description as mean ± SD and sufficient data of outcomes at both baseline and the end of intervention in each group or the net change values providedTwo investigators independently searched and assessed articles for eligibility according to the following inclusion criteria:incomplete data;duplicate publication of articles;obscurely reported outcomes, or lack of control groups; andanimal as well as non-interventional studiesStudies were excluded based on the following criteria:Two reviewing authors independently screened titles or abstracts to exclude studies that failed to meet the inclusion criteria and then obtained the full-text reports for further evaluations. Discrepancies were resolved through consensus. Detailed data of study design, context, participants’ information, interventions and outcomes were extracted. For multi-armed studies, pairs of arms relevant to the review were compared. Data of the control group were used for each intervention group comparison. The weight assigned to the control group was reduced by dividing the control group number (N) by the number of intervention groups. The selection process in sufficient detail to complete a PRISMA flow diagram was recorded in the supplementary data.In order to assess the risk of bias for each of the included studies, the two reviewing authors independently assessed the quality of the included studies according to the Cochrane risk of bias tool  and follFunnel plots were used to assess the potential existence of bias (were not shown). We performed a regression asymmetry test for detection of bias . In thisparticipants’ health status type of intervention (supplements/extract vs. food)If we identified substantial heterogeneity and there were sufficient data, we would investigate it using subgroup as well as sensitivity analyses. The following subgroup analyses were carried out:(i) Trials at high risk of bias, as specified in the Assessment of risk of bias in included studies section.(ii) Trials with small sample sizes (less than 15 participants in each group);The aim of the sensitivity analysis was to examine the effect size when some studies had less stringent inclusion criteria. If there were sufficient data to allow for sensitivity analysis, this analysis would be performed for examining the effects on results by excluding:Moreover, the leave-one-out analysis was performed to assess the key studies with substantial impact on between-study heterogeneity.Random‐effects model meta‐regression analyses were performed to assess sources of inter-trial heterogeneity. The follow-up time, ACN dose and variables at baseline (according to model and dependent variable) were considered as covariates.Meta-analysis was performed using STATA version 17.0 . For continuous outcomes, a mean difference (MD) and 95% confidence interval (CI) calculated for each study (i.e. intervention group minus control group differences).The data reported as median were converted to mean and SD using the formula recommended by the Cochrane Collaboration Handbook .2 tests. The fixed effect model was used as pooling method when there was no statistically significant heterogeneity (I2 < 50%), whereas the random-effects model was employed on the contrary (I2 > 50%).In addition, heterogeneity was assessed using Q and IThe Egger’s test was used to assess risk of publication bias.1) There was no control group;2) The studies provided no available data.Initially, 2,512 studies were found in a preliminary search from which 2,444 studies were excluded after removing duplicates and primary screening of titles and abstracts, another 21 studies were excluded due to following reasons:Finally, a total of 47 studies were selected for the analysis. A flow diagram showing the selection process is presented in Fig. The basic characteristics of the 47 randomized clinical trials (RCTs) are listed in Table A total of 23 trials compared provision of ACN supplement/extract in comparison with placebo or no intervention –63, 65.All trials provided data on losses to follow-up; eleven studies reported losses of > 10% , 55, 65 2) after intervention with ACN were evaluated in twenty studies. From two studies that had multi-arm intervention, two pairs of arms were included in the analysis [Changes of BMI ,− 0.21; 95% CI, − 0.38, − 0.04; P < 0.001)  with the difference of means as the dependent variable and the follow-up time , ACN dose  and BMI at baseline  as covariates showed no significant effects.After conducting stratified analyses according to the type of ACN (supplement/extract vs. food), we found a statistically significant effect in both subgroups Fig. . HoweverSensitivity analyses were performed and made no remarkable change in overall results . The results of leave-one-out analysis were shown in the .Eight trials compared the effect of ACN intake, as supplement/foods, versus control on waist circumference (WC). From one trial, two paired-arms were included in the analysis. Pooled data showed no statistically significant effect on WC  .The results of leave-one-out analysis were shown in the . Pooled data showed statistically significant effect on FM changes   with the difference of means as the dependent variable and the follow-up time  and used dose of ACN  as covariates showed the insignificant effects.There were no data to allow for sensitivity analysis. The Egger’s tests indicated no significant publication bias (P = 0.881).The results of leave-one-out analysis were shown in the  and health status of participants (with diabetes vs. without diabetes). Significant reductions were observed in both subgroups of T2D patients  and participants without diabetes . However, reduction of serum glucose concentrations was significantly more in the subjects with T2D (p < 0.001).After a subgroup analysis, we found that FBG decreased significantly only in those who used ACN as supplement/extract   and this association remained significant after adjusting for duration of intervention, age of participants, dose of ACN and baseline BMI .Sensitivity analyses were performed to test the robustness of the results and made no remarkable change in overall results .The results of leave-one-out analysis were shown in the .Eleven randomized controlled trials (two paired-arms from two studies) (n = 653) had measured fasting insulin. Results showed no significant effect of ACN intake on fasting insulin   and intervention type (supplement/extract vs. food). Analyses did not show any significant influence . The regression asymmetry test was not significant (p = 0.413).The results of leave-one-out analysis were shown in the supplementary file  that had examined the effects of ACN intake on HbA1c. Meta-analysis results indicated that ACN consumption decreased HbA1c . However, subgroup analysis showed this effect only in the participants with T2D  and in those trials that used ACN as supplement/extract   and this association remained significant after adjusting for ACN dose and baseline BMI .Sensitivity analyses were performed to test the robustness of the results and revealed no remarkable change in overall results . The Egger’s tests indicated no significant publication bias (P = 0.532).The results of leave-one-out analysis were shown in the  and intervention type (supplement/extract vs. food). Furthermore, there were no significant differences between subgroups  , baseline BMI  and age of subjects . Nevertheless, changes in TG were significantly associated with the baseline TG concentrations .Results of sensitivity analysis showed significant effect of ACN intake on serum TG similar to the overall effect . The regression asymmetry test was not significant (p = 0.971).The results of leave-one-out analysis were shown in the  involving 2433 subjects showed that consumption of ACN, as compared with placebo, reduces serum TC concentration   and intervention (supplement/extract vs. food). The mean change in TC concentration in trials using supplement/extract  was greater than from daily foods . Also, studies with baseline dyslipidemia showed greater decreases in TC concentrations  than those without baseline dyslipidemia  . Adjustments for ACN dose , duration of intervention  and baseline BMI  revealed no significant effect of these variables.Sensitivity analyses were performed to test the robustness of the results and showed effects similar to the overall effects . The regression asymmetry tests suggested that no significant publication bias (P = 0.200).The results of leave-one-out analysis were shown in supplementary file   , dose of intervention , BMI at baseline  and LDL-C concentrations at baseline .Results of sensitivity analysis showed significant effect of ACN intake on serum LDL-C MD:—7.72 mg/dL; 95% CI − 11.23, − 4.21 mg/dL; P < 0.001). The regression asymmetry test was not significant (p = 0.365).The results of leave-one-out analysis were shown in . Meta-analysis revealed positive effect of ACN intake on HDL-C concentrations   and intervention (supplement/extract vs. food). The mean change in serum HDL-C concentration in those trials on subjects with dyslipidemia  was greater than the trials conducted on participants normal blood lipids  , dose of intervention , BMI at baseline  and serum HDL-C concentrations at baseline .Sensitivity analyses were performed to test the robustness of the results and showed positive effect of ACN intake on HDL-C concentrations similar to the overall effects . The Egger’s tests indicated no significant publication bias (P = 0.182).The results of leave-one-out analysis were shown in  on Apo A concentrations , dose of intervention , BMI at baseline  and Apo A concentrations at baseline .Results of sensitivity analysis showed similar significant effect of ACN intake on serum apo A, . The regression asymmetry test was not significant (p = 0.301).The results of leave-one-out analysis were shown in . There was no statistically significant changes in circulating apo B in subgroup analysis, either , dose of intervention , BMI at baseline  and apo B concentrations at baseline .Results of sensitivity analysis showed no significant effect of ACN intake on serum apo B, . The regression asymmetry test was not significant, either (p = 0.128).The results of leave-one-out analysis were shown in  that is characterized by central obesity, dysglycemia, above normal blood pressure, low serum HDL-C and high TG concentrations . There iWe found the positive effect of ACN consumption on reduction of BMI which further confirms previous reports , 72. NevSensitivity analysis using the leave-one-out approach indicated that the direction of the combined estimates did not vary markedly with the removal of each study in turn. However, removal of Stull et al.  that useSome evidence indicates that truncal adiposity is the core constituent of the metabolic derrangements observed in MetS , 74. OurThe association between body FM variability and CVD risk has already been reported . NonetheThe exact mechanism of anti-adipogenesis function of ACN has not been fully elucidated yet. Some evidence suggests an inhibitory effect of ACN against adipocyte differentiation that might be exerted through 5' adenosine monophosphate-activated protein kinase (AMPK) signaling pathway activation , 81.The findings of this meta-analysis confirmed the alleviative effect of ACN on glycemic status including FBS and HbA1c. Notwithstanding, the effect on both variables was more prominent in those subjects with above-normal values, notably in the subjects with T2D. It has been documented that ACN has the potential to upregulate and also downregulate certain glucose metabolic pathways including AMPK whereby exerts its modulating effect on glucose homeostasis . Other pWe found better effect on HbA1c of ACN when consumed as a supplement. It has been shown that ACN in foods may interact with other biomolecules including proteins and polysaccharides that may affect ACN activity . EvaluatOur findings documented the effects of ACN on healthy changes of blood lipids including a decrement of TG, TC and LDL-C and an increment of HDL-C. Again here, better effects were achieved in the subjects with deranged blood lipid profile. Nevertheless, data from this meta-analysis failed to show any effect of ACN on serum concentrations of apo A and apo B. The mechanism of action of ACN on healthy modification of blood lipids is under investigation. Some studies indicate that ACN can reduce TC possibly via increased fecal excretion of neutral and acidic sterols together with down-regulation of hepatic β-hydroxy β-methyl glutaryl coenzyme A (HMGCoA) reductase, the rate-limiting enzyme of cholesterol biosynthesis . The mecThis study has some limitations. We did not evaluate the antioxidant effect of ACN that can have a major contribution to prevention of CVD , 90 and The other limitation of our study is scanty available RCTs that evaluated the long-term effects of ACN intake. Though some RCTs had treatment durations of up to two years, the question of the effectiveness of long-term ACN intake remains unanswered due to the lack of experimental evidence.In some RCTs, especially those that used ACN-containing foods, the exact amount of ACN consumption was not reported. Tt was not, therefore, possible to perform dose–response analysis. Besides, in most RCTs, ACN intake from dietary sources was not adjusted and this could affect the results.The intervention strategy across the RCTs was diverse with respect to the dosage of ACN, the frequency of intake and type of intervention (food/supplement), which inevitably contributed to the high levels of heterogeneity. We attempted to reduce the heterogeneity by choosing appropriate statistical models and using sensitivity analyses. Notwithstanding, this point should be considered when interpreting and generalizing the results.Finally, the present meta-analysis employed just the papers published in English. Thus, the papers in other languages or unpublished data might potentially increase the heterogeneity in the present results.The findings of this meta-analysis indicate that ACN intake in the forms of natural foods and supplements can induce healthy changes in body FM, glycemic and lipidemic status and these effects are more prominent in the subjects with above-normal values. Further well-designed studies are needed to evaluate and compare the efficacy of ACN consumption as natural food sources versus supplements. Meanwhile, fortification of foods and drinks with ACN can also be considered to increase ACN intake in the general population as a preventive strategy against many diseases notably CVD.Additional file 1: Figure 1S. Risk of bias summary: Reviewer authors’ judgements about each “risk of bias” item for every single included study; Figure 2S. Risk of bias graph: review authors' judgments about each risk of bias item presented as percentages across all included studies; Figure 3S. Forest plot of the effect of anthocyanins compared with control on body mass index (kg/m2); Figure 4S. The result of leave-one-out analysis of the effect of anthocyanin on body mass index (kg/m2); Figure 5S. Forest plot of the effect of anthocyanins compared with control on waist circumference (cm); Figure 6S. The result of leave-one-out analysis of the effect of anthocyanin on waist circumference (cm); Figure 7S. Forest plot of the effect of anthocyanins compared with control on body fat mass (%); Figure 8S. The result of leave-one-out analysis of the effect of anthocyanin on body fat mass (%); Figure 9S. Forest plot of the effect of anthocyanins compared with control on fasting blood glucose (mg/dL); Figure 10S. The result of leave-one-out analysis of the effect of anthocyanin on fasting blood glucose (mg/dL); Figure 11S. Forest plot of the effect of anthocyanins compared with control on fasting serum insulin (µIU/mL); Figure 12S. Forest plot of subgroup analysis of the effect of anthocyanins compared with control on fasting serum insulin (µIU/mL); Figure 13S. The result of leave-one-out analysis of the effect of anthocyanin on fasting serum insulin (µIU/mL); Figure 14S. Result of subgroup analysis of the effect of anthocyanins compared with control on HbA1c (%); Figure 15S. The result of leave-one-out analysis of the effect of anthocyanin on HbA1c (%); Figure 16S. Forest plot of the effect of anthocyanins compared with control on serum triglycerides concentrations (mg/dL); Figure 17S. Result of subgroup analysis of the effect of anthocyanins compared with control on serum triglyceride concentrations (mg/dL); Figure 18S. The result of leave-one-out analysis of the effect of anthocyanin on serum triglyceride concentrations (mg/dL); Figure 19S. Forest plot of the effect of anthocyanins compared with control on serum total cholesterol (mg/dL); Figure 20S. Result of subgroup analysis of the effect of anthocyanins compared with control on serum total cholesterol (mg/dL); Figure 21S. The result of leave-one-out analysis of the effect of anthocyanin on serum total cholesterol concentrations (mg/dL); Figure 22S. Forest plot of the effect of anthocyanins compared with control on serum LDL-C concentrations (mg/dL); Figure 23S. The result of subgroup analysis of the effect of anthocyanins compared with control on serum LDL-C concentrations (mg/dL); Figure 24S. The result of leave-one-out analysis of the effect of anthocyanin on serum LDL-C concentrations (mg/dL); Figure 25S. Forest plot of the effect of anthocyanins compared with control on serum HDL-C concentrations (mg/dL); Figure 26S. Result of subgroup analysis of the effect of anthocyanins compared with control on serum HDL-C concentrations (mg/dL); Figure 27S. The result of leave-one-out analysis of the effect of anthocyanin on serum HDL-C concentrations (mg/dL); Figure 28S. Forest plot of the effect of anthocyanins compared with control on serum Apo A concentrations (mg/dL); Figure 29S. Result of subgroup analysis of the effect of anthocyanins intake compared with control on serum Apo A concentrations (mg/dL); Figure 30S. The result of leave-one-out analysis of the effect of anthocyanin on serum Apo A concentrations (mg/dL); Figure 31S. Forest plot of the effect of anthocyanins compared with control on serum Apo B concentrations (mg/dL); Figure 32S. Result of subgroup analysis of the effect of anthocyanins intake compared with control on serum Apo B concentrations (mg/dL); Figure 33S. The result of leave-one-out analysis of the effect of anthocyanin on serum Apo B concentrations (mg/dL).


However, in meta-regression and subgroup analysis we observed significant associations between the SMD of paraoxonase and age (p = 0.003), HDL–cholesterol (p = 0.029), and study country (p = 0.04), and the SMD of arylesterase and age (p = 0.007), body mass index (p = 0.012), HDL–cholesterol (p = 0.002), and pharmacological treatment for schizophrenia (p < 0.001). In the absence of overall between-group differences, our systematic review and meta-analysis suggests that alterations in paraoxonase-1 may reflect a pro-oxidant state in specific subgroups of patients with schizophrenia that require further assessment in appropriately designed studies.The presence of a pro-oxidant state in patients with schizophrenia may account for the increased risk of atherosclerosis and cardiovascular disease in this group and supports the potential utility of circulating biomarkers of oxidative stress for risk stratification and management. We investigated this issue by conducting a systematic review and meta-analysis of the association between the circulating concentrations of paraoxonase-1, an antioxidant calcium-dependent high-density lipoprotein (HDL)-associated esterase, with paraoxonase and arylesterase activity in schizophrenia. We searched electronic databases from inception to 31 May 2023 for studies investigating paraoxonase-1 in patients with schizophrenia and healthy controls and assessed the risk of bias and the certainty of evidence (PROSPERO registration number: CRD42023435442). Thirteen studies were identified for analysis. There were no significant between-group differences in paraoxonase (standard mean difference, SMD = 0.12, 95% CI −0.23 to 0.48,  There is increasing evidence from experimental models and human studies that schizophrenia and associated psychiatric disorders, e.g., schizoaffective disorder, are characterised by a dysregulated immune system and a pro-oxidative state ,2,3,4,5.p = 0.005) [The paraoxonase multi-gene family is widely expressed in mammals and includes paraoxonase-1, paraoxonase-2, and paraoxonase-3 . Paraoxo= 0.005) . TherefoWe investigated this issue by conducting a systematic review and meta-analysis of studies reporting circulating paraoxonase-1 concentrations in patients with schizophrenia and healthy controls and assessing possible associations between the effect size of paraoxonase and arylesterase activity and pre-defined variables. We speculated that schizophrenia is associated with a significant reduction in the enzymatic activity of paraoxonase and arylesterase, reflecting the presence of a pro-oxidant state in this patient group.We searched PubMed, Web of Science, and Scopus from inception to 31 May 2023. The following terms and their combination were used in the systematic search: “paraoxonase” or “PON” or “arylesterase” and “schizophrenia” or “acute psychotic disorder” or “schizoaffective disorder”. Two independent investigators screened the abstracts and, if relevant, individual articles according to the following criteria: (i) the assessment of plasma or serum paraoxonase-1 concentrations; (ii) the assessment of paraoxonase and/or arylesterase activity in patients with schizophrenia and in healthy controls (case-control design); (iii) the assessment of 10 or more patients; (iv) the use of English language; and (v) the availability of the full-text of the article. The 2 investigators also searched the references of individual articles for additional studies. A third investigator was involved in case of disagreement.Extracted parameters included: first author, year of publication, study country, number of participants, age, sex distribution, body mass index, serum concentrations of paraoxonase, arylesterase, and HDL–cholesterol, and pharmacological treatment for schizophrenia. We assessed the risk of bias  and thep < 0.05). Means and standard deviations were extrapolated from medians and interquartile ranges or ranges [p < 0.10) and the I2 statistic. [Forest plots of standardized mean differences (SMD) and 95% confidence intervals (CIs) were generated to evaluate the presence of differences in paraoxonase-1 concentrations between patients with schizophrenia and healthy controls  and 512 healthy controls  ,69,70,71p = 0.50; I2 = 87.1%, p < 0.001; In forest plots, paraoxonase activity was non-significantly different between control subjects and patients with schizophrenia  but not with the Egger’s test (p = 0.10). The “trim-and-fill” procedure identified four studies to be added to the funnel plot (p = 0.25).There was a significant publication bias with the Begg’s test (nel plot . The resp = 0.13), the publication year , the sample size , or the body mass index  in the meta-regression analysis. By contrast, there was a significant negative association with the patient/control age ratio  in studies conducted in Turkey  than those conducted in other countries  in the pooled SMD between studies performed in untreated patients  and those in treated patients  was downgraded to extremely low  after considering the high and unexplained heterogeneity, the relatively high imprecision, and the presence of publication bias.Ten studies including 14 study groups, all with a low risk of bias , reportep = 0.61; I2 = 82.8%, p < 0.001; In forest plots, arylesterase activity was non-significantly different between control subjects and patients with schizophrenia , the Egger’s test (p = 0.75), or the “trim-and-fill” method (There was no publication bias with the Begg’s test (” method .p = 0.31), the publication year , or the sample size . By contrast, we observed significant positive associations between the effect size and the patient/control age ratio  in SMD values between studies conducted in Turkey  and those conducted in other countries  but not in those conducted in treated patients  was downgraded to very low  after taking into account the relatively high imprecision .In our systematic review and meta-analysis, the enzymatic activity of paraoxonase and arylesterase, two essential components of circulating paraoxonase-1, were non-significantly different between patients with schizophrenia and healthy controls. However, in a meta-regression analysis we identified the presence of significant associations between the effect size of paraoxonase activity and patient/control age and HDL–cholesterol ratios, and between the effect size of arylesterase activity and patient/control age, body mass index, and HDL–cholesterol ratios. Furthermore, in a subgroup analysis the effect size of paraoxonase activity was significantly associated with the study country, whereas the effect size of arylesterase was significantly associated with pharmacological treatment for schizophrenia. In sensitivity analyses, the pooled SMD values of paraoxonase and arylesterase activity were not substantially influenced by individual studies, suggesting stability of the results of the meta-analysis.The increasing evidence of a state of oxidative stress associated with schizophrenia has stimulated a significant body of research investigating the clinical significance of alterations in the redox state in this condition as well as other associated co-morbidities. In this context, studies have recently investigated the association between several biomarkers of oxidative stress, the duration of schizophrenia, and specific symptoms and psychopathological features of the disease, with contrasting results in terms of the direction of the observed associations ,74. PendIn terms of inter-individual variability in patient characteristics, there is good evidence that the redox state can be significantly influenced by age , gender p < 0.001), cerebrovascular disease , and congestive heart failure . Similar results were observed in longitudinal studies where schizophrenia was significantly associated with incident cardiovascular disease , incident coronary heart disease , incident cerebrovascular disease , and incident congestive heart failure  [p = 0.03). Interestingly, a significant increase in HDL–cholesterol was observed in patients who responded to three-week treatment with atypical antipsychotic drugs  but not with conventional antipsychotics  [The presence of a strong association between schizophrenia and atherosclerotic cardiovascular disease has been highlighted in a systematic review and meta-analysis that included 38 studies, of which 29 were longitudinal, assessing a total of 1,591,106 patients with schizophrenia. In cross-sectional studies, schizophrenia was significantly associated with the presence of coronary heart disease  . A numbe= 0.009) ,83. Stud= 0.009) ,85,86. I = 0.82) . The ben = 0.82) . Additiop = 0.014) after adjusting for age, urea, creatinine, uric acid, thyroxine, leukocytes, body mass index, and diastolic blood pressure [Another interesting observation in our subgroup analysis was the presence of significant reductions in arylesterase activity in studies of untreated patients but not in those of patients receiving pharmacological treatment for schizophrenia. Albeit hypothesis-generating, this finding suggests that treatment with antipsychotic drugs might increase arylesterase activity to levels that are comparable to those observed in healthy subjects. This would dilute the differences in arylesterase activity between treated patients and healthy controls. A systematic review and meta-analysis published in 2011 did not report any conclusive evidence regarding the effect of antipsychotic drugs on oxidative stress . Howeverpressure . Large, Strengths of our study are the combined assessment of paraoxonase and arylesterase enzymatic activity, essential components of paraoxonase-1, the comprehensive set of meta-regression and subgroup analyses, the rigorous assessment of the risk of bias and the certainty of evidence, and the stability of the results after sequentially removing individual studies in a sensitivity analysis. A significant limitation is that a significant proportion of studies was conducted in one country, Turkey, which limits the generalisability of the results to other patient populations, particularly European, North American, and Eastern Asian patients. This issue requires further research as studies conducted in USA and South Africa have reported the presence of significant ethnic-related differences in paraoxonase-1 ,101,102.Our study has shown the presence of non-significant differences in circulating paraoxonase-1, expressed as paraoxonase and arylesterase enzyme activity, between patients with schizophrenia and healthy controls. However, the results of meta-regression and subgroup analyses suggest that such differences may be significant in subgroups of patients with schizophrenia, particularly older patients with relatively low HDL–cholesterol concentrations. Further studies should address this issue by investigating paraoxonase-1 and the possible effects of antipsychotic drugs and other antioxidant strategies on disease progression and cardiovascular risk in these subgroups and consider the modulating effect of genetic polymorphisms of paraoxonase-1 and ethnic differences.


Premature ventricular complexes (PVCs) are a potentially reversible cause of heart failure. However, the characteristics of patients most likely to develop impaired left ventricular function are unclear. Hence, the objective of this study is to systematically assess risk factors for the development of PVC-induced cardiomyopathy.We performed a structured database search of the scientific literature for studies investigating risk factors for the development of PVC-induced cardiomyopathy (PVC-CM). We investigated the reporting of PVC-CM risk factors (RF) and assessed the comparative association of the different RF using random-effect meta-analysis.A total of 26 studies (9 prospective and 17 retrospective studies) involving 16,764,641 patients were analyzed . Eleven RF were suitable for quantitative analysis (≥ 3 occurrences in multivariable model assessing a binary change in left ventricular (LV) function). Among these, age , the presence of symptoms , non-sustained ventricular tachycardias (VT) , LV origin , epicardial origin , the presence of interpolation , PVC duration , and PVC burden  were all significantly associated with PVC-CM.In this meta-analysis, the most consistent risk factors for PVC-CM were age, non-sustained VT, LV, epicardial origin, interpolation, and PVC burden, whereas the presence of symptoms significantly reduced the risk. These findings help tailor stringent follow-up of patients presenting with frequent PVCs and normal LV function.The online version contains supplementary material available at 10.1007/s10840-022-01421-8. Premature ventricular complex-induced cardiomyopathy (PVC-CM) is defined as the development of left ventricular dysfunction (left ventricular ejection fraction (LVEF) of < 50%) caused solely by frequent PVCs . SuperimClinical studies have found that a high PVC burden is associated with an increased risk of systolic heart failure (HF) (hazard ratio [HR]: 1.48 to 1.8) , 4. Two Prior studies investigating risk factors for PVC-CM were retrospective and were not designed with the main objective of assessing these RFs , 11, 12.We therefore conducted a systematic review and meta-analysis of studies addressing clinical, ECG, Holter, or echocardiographic risk factors able to differentiate patients having a PVC-induced CM from other forms of CM.This systematic review and meta-analysis received approval from the ethics committee and was registered on PROSPERO (CRD42021243622). The reporting of our results was done according to the PRISMA statement about systematic reviews and meta-analyses   RCTs, prospective, or retrospective observational studies and registers; (2) with at least 50 patients total (with and without PVC-CM); (3) assessing adult patients with at least part of the cohort diagnosed with PVC-CM and at least part of the cohort presenting with PVCs; (4) investigating risk factors for the development of PVC-CM (which were not defined beforehand); (5) reporting summary statistics such as regression coefficients, odds ratios (OR) or HR; (6) assessing the incidence, prevalence, or recovery of heart failure thought to be related to PVCs or the change in ejection fraction (EF) due to the presence, increase, or reduction of PVCs; (7) providing either time-to-event data or cross-sectional data; and (8) providing at least one adjusted (multivariable) risk-factor model.The primary endpoint of this meta-analysis was the quantitative meta-analysis of risk factors for the development of PVC-CM. We pre-defined that risk factors should be reported in at least 3 different studies with a compatible definition in order to allow for a meaningful quantitative summary.Secondary endpoints were either the qualitative analysis of risk factors reported in ≥ 3 different studies or important study characteristics, such as (1) the prevalence of comprehensive work-up to ensure patients diagnosed with PVC-CM did not present with another cause for heart failure; (2) the differences in the reported definitions of PVC-CM; and (3) the assessment of study quality using the validated QUIPS  tool .The primary outcome of this meta-analysis was the presence of PVC-CM, which we pre-defined either as the development, presence, or recovery from heart failure with reduced ejection fraction (HFrEF) in patients with CMP in whom no other cause of heart failure was evident. Further details are available in the supplemental.A meta-analysis was conducted on risk factors presenting ≥ 3 times throughout the studies. When continuous risk factors were presented using cutoffs, the exposure per group (above and below the respective cutoff) was derived as recommended in previous dose-exposure meta-analyses and corresponding guidelines –19.Further details regarding the analysis of risk factors are given in the supplemental.Study quality was assessed according to the QUIPS tool  and summThe analysis was performed according to the recommendations of the Cochrane Collaboration  and the We recorded quantitative measures of baseline characteristics as mean with standard deviation (SD) or median with interquartile range (IQR). To allow for quantitative summaries, we transformed the median with IQR into mean with SDs using a mathematical transformation as proposed in previous research .For the main analysis, in order to increase the number of studies available for the quantitative summary of each risk factor, we summarized odds ratios and hazard ratios as a common measure of risk ratio, as it has been conducted in previous meta-analyses , 22.I2 statistic [To allow for the expected heterogeneity in effect measures across studies, summary relative risk estimates and their 95% CIs were estimated from a random effect model  that usetatistic .Details of the dose–response analysis are available in the supplemental.I2 statistic of > 50%.Significant heterogeneity was defined as an Evidence for publication bias was assessed for PVC burden graphically using contour-enhanced funnel plots  and the The risk of bias within each study was assessed using the QUIPS tool.P values < 0.05 were considered as significant.All statistical analyses were performed using the Statistical Software “R” . A total of 1567 studies were identified and 1540 were excluded. There were 65 full-text publications reviewed, of which 39 were excluded: 31 studies were based on the same cohorts (mostly representing abstracts of otherwise available complete studies) and 8 studies did not provide risk factors of interest or appropriate statistics. This resulted in 26 studies included in the present systematic review and meta-analysis , 28–46  studies provided a definition of PVC-CM: the CMP was mostly defined as an LVEF < 50% and 9/26 (34.6%) studies took a time component into account . The requirement for LVEF improvement in the PVC-CM definition varied from 10 to 15% in these studies.Often, several groups were analyzed in each study, which did not always report data for the overall cohort. The analyzed groups are presented in Table Most of the studies assessed the presence of PVC-CM (17/26), the recovery of LVEF after PVC-CM 4/27 (defined as a binary variable), or the worsening of LVEF suspected to be due to PVC-CM 2/27 . We conducted a pooled analysis for these three outcomes, as these are solely different ways to define a PVC-CM. Studies reporting continuous LVEF change over time 3/26) were rare , followed by sex (13/26), PVC origin (11/26), PVC and morphology (10/26), and PVC and QRS duration (each in 8/26 studies). Only few other risk factors  were investigated in ≥ 3 studies and suitable for quantitative summary. Further investigated risk factors were baseline LVEF, coupling interval, polymorphic PVCs, and outflow tract origins. These risk factors did not appear often enough (< 3 appearances) or were differently defined, hence not suitable for quantitative summary.When summarized quantitatively, age , the presence of symptoms , non-sustained VTs , LV origin , epicardial origin , the presence of interpolation , PVC burden , and PVC duration  were all significantly associated with PVC-CM , non-suQ test for residual heterogeneity was highly significant, with an I2 statistic of 89.7%.In the dose–response analysis encompassing 7 studies reporting PVC burden at different cutoffs, there was a highly significant association between increase in PVC burden and an exponential increase in risk for PVC-CM  with each increase in publication year, meaning that studies published in 2020 displayed a non-significant 2.8% lower risk association of PVC-CM with PVC burden as compared with the studies published in 2010.The PVC-CM risk associated with PVC burden decreased of 0.28%  with each increase in quality point of the summed QUIPS tool, meaning that studies with a low risk of bias (in mean 45 points in the summed QUIPS tool) presented a 2.7% higher risk association of PVC-CM with PVC burden as compared with the studies with high risk of bias (in mean 15 points in the summed QUIPS tool).Inversely, the PVC-CM risk associated with PVC burden increased of 0.09% .On funnel plot analysis of PVC burden, study distribution was mildly asymmetric Fig.  but the As presented in Fig. four major findings. First, despite screening abstracts published over 30 years of scientific research, only few studies presented a multivariable assessment of risk factors potentially associated with PVC-CM and the quality of the research currently does not allow for definitive conclusion. Second, although many candidate risk factors were proposed by the analyzed studies, only 13 risk factors  were reported often enough with appropriate statistics to allow for a quantitative summary. Many other predictors remain possible candidates for the risk stratification of PVC-CM development. Third, age, non-sustained VTs, LV and epicardial origin, interpolation, PVC duration, and PVC burden were all associated with an increased risk for PVC-CM, whereas the presence of symptoms significantly reduced the risk. Fourth, there was a clear association between increasing PVC burden and increasing PVC-CM risk. In the dose–response analysis encompassing 7 studies reporting PVC burden at different cutoffs, there was a highly significant association between increase in PVC burden and an increasing risk for PVC-CM. Specifically, per % increase in PVC burden, there was an exponential increase in the absolute risk of PVC-CM. This association was not significantly impacted by the study publication year, suggesting that despite improvements in heart failure treatments and prevention over years, the burden remains an important predictor of PVC-CM development.This systematic review and meta-analysis investigated 26 studies to investigate risk factors associated with the development of PVC-CM. We report To the best of our knowledge, this is the first systematic review and meta-analysis comprehensively assessing the risk factors for the development of PVC-induced cardiomyopathy. The optimal approach to frequent PVCs (> 10% burden) without LV dysfunction, symptoms, or idiopathic ventricular fibrillation is unclear, but patients should probably be monitored every 6–12 months with echocardiography and PVC burden assessment . TherefoSeveral studies have confirmed a correlation between a higher PVC burden and development of cardiomyopathy, although no precise burden of PVCs consistently predicts the development of a cardiomyopathy. In this meta-analysis, we found a highly significant association between an increase in PVC burden and increasing risk for PVC-CM.This systematic review and meta-analysis has several limitations. First, the quantitative summary of risk factors we are presenting summarizes different measures of risks (odds and hazard ratios) together. While this has been conducted in previous research and is acknowledged by recent guidelines as a possible necessary simplification , this miThird, as several studies did not thoroughly assess other underlying heart failure etiologies in their patients collectively, our estimates may have been occasionally confounded by other causes of heart failure. Fourth, as most of the studies providing a PVC burden cutoff only provided two categories, we had to assume a linear trend between PVC exposure and the associated increase in risk . With more detailed data, quadratic estimations could lead to more accurate dose–response relationship modelling.In this meta-analysis, the most consistent risk factors for PVC-CM were age, non-sustained VTs, LV and epicardial origin, interpolation, PVC duration, and PVC burden, while the presence of symptoms significantly reduced the risk. These findings help tailor stringent follow-up to patients presenting with frequent PVCs and normal LV function.Supplementary file1 (DOCX 861 KB)Below is the link to the electronic supplementary material.


Trichomonas vaginalis (TV) infection and cervical neoplasia is conflicting. The main aim of this study was to evaluate the magnitude of the risk of cervical neoplasia associated with TV infection.The evidence in the literature regarding the relationship between TV infection with cervical neoplasia, was performed. For this aim, we searched scientific databases  from inception to March 15, 2023. A random-effects model was applied by Stata 17.0 to calculate the pooled and adjusted odds ratios (ORs) with 95% confidence intervals (CI), including subgroup, sensitivity, and cumulative analyses to explore sources of heterogeneity.A meta-analysis of observational studies, which provided raw data on the association of I2 = 87.7%) and adjusted  ORs indicated a significant positive association between TV infection and the development of cervical neoplasia. There was no significant change in pooled and adjusted ORs by applying sensitivity and cumulative analyses, indicating the robustness of our findings. The pooled OR was significant in most sub-group analyses. There was no publication bias in the included studies.Of the 2584 records initially identified, 35 eligible studies contributed data for 67,856 women with cervical neoplasia, and 933,697 healthy controls from 14 countries were included. The pooled (2.15; 1.61–2.87; TV infection are at significantly greater risk of cervical neoplasia. Future research, particularly longitudinal and experimental studies, should be done to better understand the various aspects of this association.Our findings indicated that women with a  Cervical cancer is one of the leading causes of cancer deaths among women worldwide . Based oTrichomonas vaginalis (TV) infections may increase the risk of cervical cancer in infected women [Even though HPV is an essential and recognized factor in abnormalities and cancer of the cervix, it is not a sufficient cause. There are several risk factors associated with HPV-associated cervical cancer, such as age at first sexual contact, the number of partners, smoking, oral contraceptives, the presence of human immunodeficiency virus (HIV), Herpes simplex, and co-infection with other genital or sexually transmitted infections (STIs) . It is ued women –15.TV is capable of infecting both the male and female reproductive systems [TV infection [TV infection and the development of cervical cancer is still not entirely understood, it is thought that the inflammatory response is the key factor [TV infection could overexpress viral tumorigenesis, which helps activate cancer-causing pathways [As a flagellated protozoan parasite,  systems . Approxi systems . The mos systems . Vaginalnfection , 20. Alty factor , 22. Infpathways .TV infection and cervical neoplasia [TV and the development of cervical cancer [TV infection and cervical neoplasia, we updated the previous meta-analysis.There are several clinical and observational studies evaluating the association between eoplasia –29. A sieoplasia ; moreovel cancer . In thisThe Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA)  and the TV infection and cervical neoplasia from database inception to March 15, 2023. We did not use any language, geographic, or date restrictions. For database searches, the following Medical Subject Headings (MeSH) terms and keywords were combined using Boolean operators : "Trichomonas vaginalis", "Trichomonas vaginitis", "Trichomonas infections", "Trichomoniasis", "Trichomonas", "sexually transmitted infection", "cervical cancer", "cervical carcinoma", "cervical neoplasm", "cervical abnormality", "cervical dysplasia", "cancer of the cervix", "carcinoma of the cervix", and "cervical intraepithelial neoplasia" (Two researchers (A.F.N. and K.B.) did a comprehensive systematic search in international scientific databases, including PubMed/Medline, Scopus, Web of Science, and Embase, for all peer-reviewed, published studies assessing the relationship between oplasia" . AdditioTV infection and cervical neoplasia using a cross-sectional, cohort, or case-control design; gave a clear explanation of the methods for diagnosis of TV infection and cervical neoplasia; and had sufficient data to calculate the odds ratio (OR) or relative risk (RR) with 95% a confidence interval (95% CI). Studies that evaluated the association following the treatment, conference abstracts, reviews, systematic reviews, editorials, case reports, case series, and papers with not available full-text were excluded. In this meta-analysis, if multiple studies reported on the same population or datasets, the most recent or comprehensive study was chosen for inclusion.After deleting duplicates, three trained researchers  independently screened the titles and abstracts and also evaluated the full-text of articles for eligibility. Ambiguous articles were examined by the lead investigator (AR). All the inconsistencies were resolved through discussion. Studies were eligible for inclusion if they investigated the correlation between TV infection, number of women with cervical neoplasia as cases, number of women without cervical neoplasia as control group, number of infected women in case and control groups. Where available, adjusted ORs were extracted and used for meta-analysis to reduce confounding, as recommended by the Cochrane Handbook [Three independent investigators  performed data extraction from each relevant study. Consulting with the lead investigator (A. R.) helped resolve any disputes. The following information was extracted (if available): first author’s name, publication year, country, geographic region, type of study, diagnostic method for Handbook . AccordiHandbook –37.P-values were determined, with P-values less than 0.05 considered statistically significant. Summary pooled and adjusted ORs and related 95% confidence intervals (CIs) were calculated using the DerSimonian-Laird random-effects model (REM) for combining results across studies, which incorporates between- and within-study variance [I2 statistics were applied to measure between-study heterogeneity. An I2 value more than 75% was considered significant heterogeneity [TV infection (polymerase chain reaction (PCR), cytology, microscopy, and other), WHO-defined geographical region , country’s income level , and risk of bias for studies (low and moderate). Sensitivity analyses were performed by excluding individual studies to examine the robustness of the results and the effect of each study on the estimated ORs. Cumulative meta-analyses were applied to determine the reliability of the estimated ORs. Publication bias was evaluated using the Egger test and visual inspection of funnel plots [Meta-analysis was performed with Stata software version 17 . For all tests, two-sided variance . We usedogeneity . We alsoel plots .TV infection was assessed by cytology in 13 studies [As shown in the PRISMA flow diagram , our ini studies , 62, 67, studies , 59, 66, studies , 60, 63, studies , 64, 65.TV infection was significantly associated with an increased risk of cervical neoplasia. A substantial heterogeneity  was observed between the studies. The sensitivity analysis indicated that the exclusion of any of the studies did not have a significant impact on the pooled OR , LGSIL/CIN1  and invasive cancer .As shown in I2 = 67.9%) and cohort  studies, while the association was not significant in cross-sectional studies . With regard to geographical regions, studies from regions of the Middle East and Africa , Europe and north America , Western pacific , South east Asia  demonstrated significant association, while studies from South America showed non-significant results . According to diagnostic methods for detection of TV infection, subgroup analysis indicated significant association in studies that used cytological , microscopic , and other  , while studies that used PCR showed a non-significant association . The association was significant in all subgroup analyses based on publication year, income levels and risk of biases. More details on subgroup analyses are presented in Subgroup analysis considering type of studies indicated significant positive associations in case-control  and adjusted ORs (I2 = 31%). Variation in diagnostic methods for the diagnosis of TV infection, definition and stages of cervical neoplasia, geographical region, country’s income levels, and publication year are likely to have contributed to the heterogeneity Click here for additional data file.S2 Fig(TIF)Click here for additional data file.S3 Fig(TIF)Click here for additional data file.S4 Fig(TIF)Click here for additional data file.S5 Fig(TIF)Click here for additional data file.S6 Fig(TIF)Click here for additional data file.S7 Fig(TIF)Click here for additional data file.S1 Checklist(DOCX)Click here for additional data file.


TNF-α 308G > A.Migraine is a complex neurological disorder that is characterized by a "lower threshold of neuronal hyperexcitability" with distinctive periodicity and complex vascular dysfunction. Genetic factors have impacted incredibly on the susceptibility of migraine and one such example is the TNF-α 308G > A risk on the susceptibility of migraine.Therefore, we aim to provide a glimpse of the association of the 2 statistics and Begg's and Egger's tests were used for finding the publication bias, tests were two-sided, and a p-value of < 0.05 was considered statistically significant. The Trial Sequential Analysis with Meta-regression Analysis were also utilized to find out the sample size requirement for meta-analysis to avoid type I error and source of heterogeneity respectively.The pooled odds ratio with the associated 95% of confidence interval were calculated using different genetic models. Heterogeneity was accessed by using Cochran's Q Test and IA total of 13 studies with cases: 7193 and controls: 23,091 were included and after using different genetic models, no overall association with migraine and its clinical subtype migraine with aura was observed  respectively. Interestingly, after sub-grouping using the “ethnicity criteria” in the migraine group, it was observed that the allelic genetic model and the dominant model were found to be significantly associated with the Asian ethnic group (OR: 1.79, 95% C.I. [1.13–2.84], and OR: 1.85, 95% C.I. .In conclusion, the present meta-analysis has provided evidence that 308G > A increases the risk of migraine only in the Asian population. Lower neuronal hyperexcitability is used to define migraine, which is a complex neurological disorder characterized by distinctive periodicity and complex vascular dysfunction . AccordiTNF-α is an astonishing example of a proinflammatory cytokine involved in inflammation initiation and is secreted by the microglial cell upon activation . The impact of the mutant allele was seen after using the random effect model as the “test of heterogeneity (I2) was 88% , a non-significant strength of association was found OR: 1.29, 95% C.I. [0.95–1.75]. As the value for the test of heterogeneity was found below the moderate threshold (55%), a fixed effect model was then utilized to check out the effect of the recessive genetic model on migraine susceptibility and found an OR: 1.00, 95% C.I. [0.84–1.19]. For the analysis of the over-dominant genetic model with an I2 = 84%, the random effect model showed the associated value OR: 1.22 [0.91–1.64].This currently updated meta-analysis utilizes the different genetic models to observe the effect of the rs1800629 variant of 88% Fig. A. After Using Egger's test, which is based on the relationship between standard error and strength of association, the publication bias was evaluated. The plotted articles with high accuracy on top and low accuracy down in the plot formed the perfect funnel-shaped structures. Such symmetrical funnel plots were formed for all genetic models which direct that there was no publication bias Fig. B. A sensAfter sub-grouping using the criteria of ethnicity for the migraine group, it was observed that the allelic genetic model and the dominant model were found to be significantly associated with the Asian ethnic group  and the random model respectively. Sub-grouping of MA based on the ethnicity criteria revealed no association between rs1800629 and migraines in Asian and Caucasian populations   strong linear relationship was observed (R = 0.6184)  . The expThe present meta-analysis did not show any significant association between the rs1800629 and migraine and its clinical phenotype (MA) after utilizing different genetic models. Interestingly, after sub-grouping using the “ethnicity criteria” in the migraine group, it was found that the dominant model showed a slightly significantly higher risk of migraine than the allelic model Table  in the AComparing our meta-analytic data with the pre-existing meta-analysis, the results were found consistent –21 in coThere are many reasons why a lack of association occurs and the prime reason might be the limited number of studies in the meta-analysis as shown Fig. , also thRegarding the strength of the present meta-analysis from the previous meta-analysis –21 firstIn the current meta-analysis, no overall association was found between the risk variant and susceptibility to migraine and its clinical type i.e., MA but after subgrouping, the increased risk was observed between -308 G/A & migraine in Asian population , 24.