CELSR2, PSRC1, and SORT1 has been associated with cardiovascular diseases. Thus, the aim of this study was (i) to perform a systematic review and updated meta-analysis of the association of three polymorphisms  of this cluster with cardiovascular diseases, and (ii) to explore by PheWAS signals of the three SNPs in cardiovascular diseases and to evaluate the effect of rs599839 with tissue expression by in silico tools. Three electronic databases were searched to identify eligible studies. The meta-analysis showed that the rs599839 , rs646776  polymorphisms showed an increased risk for cardiovascular diseases. PheWas analysis showed associations with coronary artery disease and total cholesterol. Our results suggest a possible involvement of the CELSR2-PSRC1-SORT1 cluster variants in the risk association of cardiovascular diseases, particularly coronary artery disease.A cluster of three genes  Nowadays, cardiovascular diseases (CVDs) are defined as the “pathological processes involving the cardiovascular system such as the heart, the blood vessels, or the pericardium” by the Medical Subject Headings criteria (MeSH). The American Heart Association, until 2020, reported that cardiovascular diseases represented the leading cause of mortality worldwide, with approximately 18.6 million deaths and 523 million cases since records began [Among the cardiovascular diseases of atherosclerotic origin is mainly coronary artery disease . TherefoCELSR2-PSRC1-SORT1 that could participate in mechanisms that involve the alteration of serum lipids and plasmatic cholesterol levels and increase susceptibility for the development of CVD [Since 2007, various genome-wide association studies (GWAS) have identified risk loci and candidate genes involved in the development of CVDs [t of CVD ,9.CELSR2 (Cadherin EGF LAG Seven-Pass G-Type Receptor 2) is a member of the flamingo subfamily, part of the cadherin superfamily. This gene encodes a cadherin responsible for contact-mediated cell adhesion and ligand-receptor cell interactions [PSRC1 (proline/serine-rich coiled-coil 1) gene is partly responsible for microtubule destabilization [SORT1 (sortilin) is a gene that encodes the sortilin 1 protein, implicated in several lipid-associated functions such as VLDL secretion, atherosclerotic lesions development, LDL-cholesterol metabolism, and PCSK9 secretion [SORT1-associated functions lead to an increased risk of cardiovascular diseases. However, the results regarding these three genes and their participation in CVDs are so far contradictory and not well understood [The ractions . The PSRlization ,12. The ecretion . It alsoecretion . Some ofderstood ,15,16. LCELSR2-PSRC1-SORT1  with cardiovascular diseases. Finally, we explored prior genome-wide association signals of the rs599839 polymorphism with other phenotypes related to cardiovascular diseases and evaluated the effect of rs599839 on gene expression in the liver using in silico tools.Therefore, this study aimed to perform a systematic review and updated meta-analysis of the association of The present work followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines  Supplem. This stA systematic literature search was performed up to December 2021 to identify relevant articles published on three widely used electronic literature databases . First, the following keywords were combined to obtain the eligible articles: (i) “rs646776”, “rs599839”, or “rs464218” and (ii) “coronary artery disease”, “cardiovascular disease”, “coronary heart disease”, “cerebrovascular disease”, “myocardial infarction”, “peripheral arterial”, “atherosclerosis”, or “ischemic stroke”. To perform a more detailed search, we also included some cardiovascular risk factors: (i) “rs646776”, “rs599839”, or “rs464218” and (ii) “hypertension”, “lipids”, “metabolic syndrome”, “glucose”, “insulin”, “type 2 diabetes mellitus”, “abdominal obesity”, or “dyslipidemia” in combination with the gene polymorphisms previously mentioned . CELSR2-PSRC1-SORT1  and cardiovascular disease susceptibility; (4) studies in which genotypes in cases and controls were available or (5) had sufficient data to calculate an odds ratio (OR) with a 95% confidence interval (95% CI); and (6) peer-reviewed studies written in the English language.Only studies meeting the following inclusive selection criteria were eligible: (1) studies that evaluated human subjects; (2) case-control study designs; (3) studies investigating the association between the genetic variants of p-value < 0.05.The exclusion criteria were meta-analyses, reviews, case reports, abstracts, editorial articles, or duplicate publications with overlapping data. For meta-analyses only, we excluded studies in which the control group had a significant Hardy–Weinberg equilibrium (HWE) Fourteen studies were eligible for systematic review and twelve for meta-analysis following the abovementioned search. The quality of the articles included in our study was evaluated using the NOS  scale. The evaluation was based on three aspects: the selection , the comparability between groups , and exposure factors , with a total score of 9 points. Studies with a score ≥6 were considered high quality for inclusion in this analysis.Two authors (RGCA and TBGC) searched, applied the eligibility criteria and quality assessment, and extracted data from eligible studies. Two other investigators (JMRP and NPH) resolved discrepancies by consensus. last name of the leading author;year of publication;self-reported ancestry and country of study population;number of cases and controls;diagnostic criteria of cases and controls;allele and genotype frequencies.The information obtained was: CELSR2-PSRC1-SORT1  and the risk of cardiovascular diseases. Statistical significance was established at a Z-test p-value < 0.05. We used the main models of inheritance reported in statistical analyses of genetic polymorphisms in epidemiological studies [1. AllelicAllele 1 (Reference)Allele 2;2. Codominant11 (Reference)1222;3. Dominant11 (Reference)12 + 22;4. Recessive11 + 12 (References)22.We calculated the odds ratio and 95% CI to estimate the association between genetic variants of p-value < 0.1 or I2 > 50% indicated heterogeneity. We used the random effects model unless otherwise stated, in which case we used the fixed effects model. For a better understanding of CVDs, we performed sub-analyses based on:overall, coronary artery disease, myocardial infarction, acute coronary artery syndrome, ischemic stroke, and peripheral arterial disease;studies including only coronary artery disease;studies including only healthy controls;Asian populations.We evaluated statistical heterogeneity between studies by a chi-square-based Cochran Q test. A Higgins I-squared statistic p-value < 0.05 suggests bias) identified any bias. Chi-squared statistics evaluated Hardy–Weinberg equilibrium (HWE) in each study. All data were analyzed using Comprehensive Meta-Analysis Software version 2. In addition, we duplicated the searches in the different databases: (a) manually by the assigned researchers , and (b) by using the COVIDENCE software (https://www.covidence.org/) (accessed on 29 December 2021) to verify searches and articles found. The p-value adjusted by the Bonferroni method was considered, and significance was set at p-value < 0.004 , and the results were expressed in scientific notation.We replicated the analyses, excluding one study at a time as a sensitivity analysis for the stability of the pooled results. We visualized publication bias by funnel plots, and the quantitative method of Begg/Egger linear regression  were identified. The flow chart for the study selection is shown in All the included studies were of quality on the NOS scale (score > 6).p-value < 0.05 in HWE. Subsequently, twelve studies were taken into consideration for further analysis. The quantitative analysis excluded two studies, by Ellis, K.L., 2011 and by Rizk, N. M., 2015, because the control group had a p-value 1 × 10−4, Q test p-value: 0.175, I2: 30.38), dominant , and recessive , p-value 1 × 10−4, Q test p-value: 0.423, I2: 1.30)  .p-value 1 × 10−4, Q test p-value: 0.265, I2: 21.54), codominant , dominant , and recessive   .p-value 1 × 10−4, Q test p-value: 0.494, I2: 0.000), and recessive   .p-value 0.290, Q test p-value: 0.040, I2: 68.81). Then, in another sub-analysis, we evaluated studies performed in the Asian population. In this sense, the outcomes evidenced a statistical association in the following model: allelic   .p-value 9 × 10−3, Q test p-value: 8 × 10−3, I2: 79.47)  . HoweverFunnel plots visualized publication bias; nonetheless, no evidence of publication bias was found in the pooled analyses. On the other hand, a sensitivity analysis was conducted to assess the influence of one study on the pooled ORs value  and whether the results can be reverted by eliminating the individual study. The result did not change after the leaving-one-out analysis, indicating the stability of the results of this meta-analysis.We performed the bioinformatic analysis only on rs599839, because it showed the highest number of significant associations in the cluster. The results from the PheWAS indicate that this genetic variant had been associated with intermediate phenotypes for CAD, total cholesterol, and LDL levels . AdditioCELSR2-PSRC1-SORT1 gene cluster and are located on chromosome 1 could be involved in the development of cardiovascular diseases through mechanisms related to serum lipid levels and plasma cholesterol levels [Cardiovascular diseases have a complex multifactorial etiology involving various risk factors that contribute to their development, including a genetic component. In this regard, several studies have suggested that genes that together form the l levels ,15.CELSR2-PSRC1-SORT1 cluster in 25224 cases and 33181 controls, investigating the association between these genetic variants and susceptibility to develop cardiovascular diseases through a systematic review and updated meta-analysis. Therefore, we analyzed three polymorphisms  in the First, after discarding heterogeneity in the included studies, we observed that the rs599839 could increase the risk of developing any cardiovascular disease  by 1.19 to 1.23 times in the allelic, dominant, and recessive models.Also, when we analyzed the rs599839 polymorphism and coronary artery disease only, we found a strong risk association that ranged from 1.17 to 1.39 times more presenting coronary artery disease in all genetic models evaluated: allelic, codominant, dominant, and recessive. These results suggest that the rs599839 polymorphism may be associated with increased susceptibility to developing coronary artery disease. Previous studies have explored the relationship between the rs599839 polymorphism and coronary artery disease ,28. AlthPSRC1, FDR = 0.0118) and myocardial infarction in the Asian population [PSRC1 gene could participate in the development of cardiovascular diseases and mainly in the genetic predisposition to coronary artery disease.Finally, we carried out a specific analysis of Asian populations. When we ruled out heterogeneity, we found that carriers of the allelic and recessive models had 1.27 and 1.30 more times, respectively, the risk of developing cardiovascular diseases. In this regard, Matsuoka, R., 2015 found an association between the rs599839 polymorphism  under the recessive model. There are few studies described in the literature on this polymorphism, and the results are controversial. Therefore, additional studies are required to elucidate the participation of this genetic variant in cardiovascular diseases [p-value, the significance does not remain. Therefore, it is possible that the results are biased. Furthermore, this sub-analysis only included the Asian population due to the lack of studies to include other populations. Moreover, we analyzed the rs464218 polymorphism of the diseases ,22. AlthCELSR2-PSRC1-SORT1 cluster, and we found an association with different cardiovascular diseases. Second, this work shows a strong risk association between the rs599839 polymorphism and coronary artery disease. Third, in this meta-analysis, we included studies in the main electronic databases and multiple searches to gather as much information as possible. On the other hand, our work has important strengths: this is the first meta-analysis involving three genetic variants  contained in the p-value was adjusted by Bonferroni correction. However, this correction could be improved in future studies with more diverse and larger sample sizes, considering other approaches applied to candidate gene studies. Fifth, we did not perform a haplotype analysis due to the few studies included in this meta-analysis. Sixth, we suggest other computational tools and approaches that could be more stringent in calculating statistical significance for future studies. We recognize that we have limitations in our study: First, due to the few studies reported in different populations, we were only able to perform a sub-analysis in the Asian population. Second, due to the small sample size for some subgroup analyses, it will be necessary to conduct additional studies with larger samples that include different populations in order to improve the reliability and stability of the meta-analysis. Third, because only studies published in the English language were included, there could be a language bias. Fourth, despite the few studies included in the present meta-analysis, the CELSR2-PSRC1-SORT1 cluster variants with an increased risk of cardiovascular diseases, particularly coronary artery disease. However, our findings warrant replication in larger sample sizes due to several limitations.In conclusion, our results suggest an association of the


Chronic kidney disease (CKD) risk is inconsistent in the normal-weight, overweight, and obese individuals due to the heterogeneity of metabolic status. This meta-analysis aimed to examine the combined effects of body mass index (BMI) and metabolic status on CKD risk.2 and/or microalbuminuria or proteinuria) associated with the body size phenotypes which are known as metabolically unhealthy normal-weight (MUNW), metabolically healthy overweight (MHOW), metabolically unhealthy overweight, metabolically healthy obese (MHO) and metabolically unhealthy obese (MUHO). The classification of subjects in included studies as metabolically unhealthy was based on the presence of three components of metabolic syndrome. BMI categorization was based on the criteria of included studies. The risk estimates and 95% confidence intervals (CIs) were extracted and pooled using random effects analysis. The MEDLINE, EMBASE, and Web of Knowledge databases were systematically searched up to March 2019 to identify all eligible studies investigating the CKD risk  in MUNW and 1.55 (95% CI = 1.34-1.79) in MHO persons. Also, MHOW was at increased risk for CKD . MUHO individuals were at the highest risk for the development of CKD . Individuals with metabolic abnormality, although at normal-weight, have an increased risk for CKD. Healthy overweight and obese individuals had higher risk; refuting the notion that metabolically healthy overweight and obese phenotypes are benign conditions. Therefo2), but express metabolic abnormalities like an increased levels of adiposity and insulin resistance and a higher susceptibility to type 2 diabetes and cardiovascular diseases (CVD) (One of the major risk factors for CKD is obesity  and obeses (CVD) ,26. MoreWhether different phenotypes of obesity have different effects on the risk of CKD is still debatable. To date, observational studies have shown inconsistent effects of obesity phenotypes on the risk of developing CKD. A single study might have low statistical power due to small sample size and other limitations. Thus, the current meta-analysis was conducted to assess the relation of different phenotypes of body size to CKD risk.Our meta-analysis was conducted according to the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines .We searched for all published observational studies that described the associations of different phenotypes of body size with the risk of CKD incident. A systematic literature search was performed using the MEDLINE, EMBASE, and Web Of Knowledge databases and was supplemented with the manual review of the reference list of obtained articles up to March, 2019. The following terms were used: ((((((((“Obesity”[Mesh] (28) OR “Body Mass Index”[Mesh]) OR body mass index[Title/Abstract]) OR obesity[Title/Abstract]) OR obese[Title/Abstract]) OR overweight[Title/Abstract]) OR normal weight[Title/Abstract]) AND ) OR  OR benign[Title/Abstract])) AND  OR chronic kidney disease[Title/Abstract]) OR CKD[Title/Abstract]) OR chronic renal disease[Title/Abstract]) OR chronic kidney insufficiency[Title/Abstract]) OR chronic renal failure[Title/Abstract]). To find studies investigating the combined effect of BMI and metabolic status on the risk of CKD and remove irrelevant studies, a specific search strategy was applied. No language restriction was applied for searching and study inclusion.Studies were considered eligible for meta-analysis if they met the following criteria: the study had a prospective cohort design; the exposures were the metabolically healthy and metabolically unhealthy phenotypes of body size and outcomes was incident of CKD; stratified subjects according to BMI categories; had one reference group in the normal-weight healthy range; reported hazard ratios (HRs), relative risks (RRs) or odds ratios (ORs) for incident of CKD; CKD incident diagnosis; and reported the criteria used for defining subject as metabolically healthy or metabolically unhealthy. Literature reviews, cross-sectional studies, case reports, republished data, and animal studies were excluded. The application of these criteria yielded 6 studies eligible for the meta-analysis.2 and/or microalbuminuria or proteinuria during at least 3 months of follow-up. The classification of subjects in included studies as metabolically unhealthy was based on the presence of at least three components of metabolic syndrome. BMI categorization was based on the criteria of included studiesThe following information was collected from each study: the author’s name, publication year, country of origin, study design, ethnicity, gender, mean or range of age, sample size, duration of the follow-up, numbers of CKD cases, definition of metabolically healthy and metabolically unhealthy phenotypes, confounding factors that were adjusted for in the multivariable analysis, and hazard ratios (HRs), relative risks (RRs) or odds ratios (ORs) with 95% confidence intervals (CIs). Data were extracted independently by two investigators, and the disagreements during the data extraction were resolved by discussion among all reviewers. CKD diagnosis as outcome was based on GFR < 60 mL/min per 1.73 mThe quality of the studies was independently assessed by the same two reviewers and any disagreement was resolved by discussion between the two investigators. The Newcastle–Ottawa Scale (NOS) was applied to assess the quality of included studies . This quThe classification of subjects in included studies as metabolically unhealthy was based on the presence of at least 3 components of metabolic syndrome by criteria from the National Cholesterol Education Program’s Adult Treatment Panel III (fasting triglyceride level ≥ 1.69 mmol/L (150 mg/dL); HDL-C < 1.04 mmol/L (40 mg/dL) in men or < 1.29 mmol/L (50 mg/dL) in women or lipid-lowering medication use; fasting glucose level ≥ 5.6 mmol/L (100 mg/dL) or antidiabetic medication use; systolic blood pressure 130 mmHg, diastolic blood pressure 85 mmHg, or use of antihypertensive medication; waist circumference > 88 cm for women and > 102 cm for men) . The sub2 statistics  (2> 50% (p < 0.1). All analyses were performed using the random-effects model . All but 1 study reported a significant difference between the groups. There was no between-study heterogeneity , with no evidence of small-study bias on the Egger’s regression test . Furthermore, compared to the MHNW phenotype, MUHOW individuals showed no significant association with risk of CKD  , with no evidence of significant heterogeneity   .Supplemental file 4). There was no significant evidence for heterogeneity in the effect sizes for these associations across studies , and no evidence of small-study bias was detected . In comparison, the corresponding pooled RR in the MUHO individuals was 2.13 (95% CI = 1.66-2.72) . However, there was significant heterogeneity in the individual estimates when the magnitude of the association was evaluated , with no evidence of small-study bias according to the Egger’s regression test   .In the present meta-analysis of 9 prospective cohort studies with 128773 participants, we addressed the combined effect of BMI and metabolic health status on the risk of CKD. The results revealed that compared with MHNW subjects, the overweight and obese individuals even in the absence of overt metabolic abnormalities, had significantly increased the risk for CKD.The differentiation of metabolically healthy overweight/obesity and metabolically impaired overweight/obesity has been suggested to have important implications for therapeutic medical decision-making ,46. As tThe findings of this meta-analysis will help establish whether MUNW and seemingly healthy overweight and obese populations face an increased risk for CKD. The main strength of the current study is its relatively large pooled sample size, enabling the determination of robust estimates for the relationship between the 6 BMI– metabolic categories and related disease that could not be estimated precisely in individual studies. Moreover, the included studies were all prospective cohort in design, which minimized the possibility of selection bias and recall bias. However, some limitations of our study should be discussed. First, because of the limited available published literature, the number of studies included in this meta-analysis is relatively small. Second, the duration of exposure to the current metabolic phenotype and longitudinal changes in metabolic status and BMI were not reported in the studies, which could partially affect the estimated risk. Studies have shown that metabolic health might be a transient condition among obese people  and appeIn conclusion, this meta-analysis of prospective studies supports the concept of heterogeneity of metabolic status among individuals within a similar BMI category. Our analysis indicated that MHOW and MHO phenotypes are positively in relation to CKD risk. The term ‘healthy’ might not be appropriate to describe these individuals because of the likelihood of underestimating the long-term effects of obesity, as well as less attention being paid to weight management. In addition, metabolically unhealthy phenotypes, including unhealthy normal weight, and unhealthy obese individuals had increased risk for CKD events, compared with MHNW individuals. Therefore, it is essential to consider both BMI and metabolic factors to reliably estimate the risk of incident CKD.


Contrary to North America and Europe, the prevalence of hypertension is rising in West Africa. Although diet is implicated as a contributor to this trend, nutritional guidelines in West Africa are not tailored to address this concern. This study aimed to address this limitation by investigating dietary factors common to West Africa and evaluating their association with hypertension.PubMed, Scopus, Web of Science, and Medline were searched to identify studies that investigated diet and hypertension in West African adults. All meta-analyses used a generic inverse-variance random effects model, with subgroup analyses by age, BMI, and study location, and were performed in R.n = 48 809 participants) satisfied inclusion criteria – all cross-sectional. Meta-analyses of the association between dietary factors and hypertension included dietary fat , red meat , junk-food , dietary salt , alcohol , and ‘fruits and vegetables’ . Subgroup analyses suggested that ‘fruit and vegetable’ consumption is less protective in the elderly.Three thousand, two hundred ninety-eight studies were identified, of which 31 (High consumption of dietary salt, red meat, dietary fat, junk food, and alcohol are associated with increased odds of hypertension, whereas high fruit and vegetable appear protective. This region-specific evidence will support the development of nutritional assessment tools for clinicians, patients, and researchers aiming to reduce hypertension in West Africa.  Hypertension, defined as a sustained elevation of blood pressure at least 140/ 90 mmHg, remains the leading preventable risk factor for cardiovascular disease and the number one cause of death globally ,3. The mThis increased trend in hypertension in Nigeria and other West African countries has been partly attributed to increased unhealthy dietary practices  and lack of physical activity ,12–15. ITo address this, an understanding of the dietary habits and risk of hypertension in Nigeria and other West African countries is needed to confirm current recommendations and improve current dietary assessment strategies to combat hypertension. Therefore, this study aimed to undertake a systematic review and meta-analysis to provide regionally specific evidence relating to the association between dietary factors and hypertension risk in West Africa that can be used to inform the development of nutritional assessment tools for clinicians, patients, and researchers and contribute towards reducing the prevalence of hypertension in West Africa.First, we performed a pilot study and evaluated the summaries of the articles we identified through our search on PubMed and which fulfilled the eligibility criteria. This step enabled us to obtain an outline of the key aspects that ought to be encompassed in the review and helped us to prepare the protocol for the systematic review and meta-analysis.The strategy for the systematic review and meta-analysis was predefined with PROSPERO CRD42022339736), which followed the guidelines for the Meta-analyses of observational epidemiological studies 22339736,.The systematic literature search was performed using a structured PICO  framework and Medical Subject Headings (MESH) indexing with key terms (and synonyms thereof). The PICO elements included the targeted West African nations (P), high dietary exposure (I), low dietary exposure (C), and hypertension/high blood pressure (O). A comprehensive search was conducted across PubMed, Ovid Medline, Scopus, and Web of Science, to identify studies published between January 2000 and February 2023. The full search strategy used for PubMed is given in Table S1 in the supplementary material. To ensure that no relevant studies were missed, the citation lists of included studies were searched to identify additional relevant studies that meet the inclusion criteria.Four of the most prevalent adverse outcomes of blood pressure were chosen as outcomes of interest for this review and meta-analysis. These included high SBP, high DBP, high mean arterial blood pressure, and high pulse pressure.P-value, means with standard deviation or standard error] that addressed any dietary intervention on hypertension or high blood pressure in healthy West African populations with participants aged at least 18 years were included in the study. Studies were excluded if the data were hypertensive individuals alone, animal and cells studies, non-West African populations, noninvestigative studies , pregnancy studies, cardiovascular diseases  or studies in cancer populations, studies that are not related to diet, food or nutrition, studies involving only alcohol, smoking, or biomarkers of diet. The screening of ‘title and abstract’ and ‘full text’ was performed in duplicate in DistillerSR   reported on the association between alcohol and odds of hypertension were included in the meta-analysis, with effect sizes ranging from 0.82 to 2.03. Compared with low consumers, the meta-analysis demonstrated that high consumption of alcohol increased the odds of hypertension by 17%  , and rank correlation test (P = 0.17) in the meta-analysis. Subgroup analysis and meta-regression did not suggest a moderating effect on the mean population age, BMI, or study location .A total of eight cross-sectional studies  reported a decrease in hypertension risk for high vegetable  and fruit intake  based on even cohort studies with 94 772 incident cases of hypertension [et al. [64] found a decreased hypertension risk based on six cohort studies  [For single ‘healthy’ food available for analysis, fruit and vegetables, we report that high fruit and vegetable intake is associated with decreased odds of hypertension. This is consistent with previous meta-analyses conducted in other regions of the world, such as those conducted by Wu et al. and Schwl.[et al.. Schwing.[et al.  reported [et al.  found a 79–0.95) –76.Numerous food groups were identified that have previously been negatively associated with hypertension : junk food, red meat, dietary fats, dietary salt, and alcohol.Ptrend = 0.045) and  women [On the basis of six cross-sectional studies with 2321 participants, we report that high consumption of junk food is associated with increased odds of hypertension in West African adults. This finding agrees with biochemical evidence that has linked junk foods, which often contain high amounts of saturated and trans fats, sodium, and by-products of oxidation, which can induce increased blood pressure in different mechanisms –83. In a0) women , while a0) women . The absI2 = 84%) [et al.[I2 = 75%). Our meta-analysis observed less heterogeneity with the confounders examined not significantly modifying the observed association. The biological effect of high meat consumption on hypertension is supported by the components found in meat, such as saturated fat, cholesterol, high sodium levels in processed meat, haem iron, and substances formed during cooking or processing of meat, such as heterocyclic amines, advanced glycosylation end-products, acrylamides, and trimethylamine-N-oxide [Our meta-analysis contributes to the growing body of evidence on the association between red meat consumption and hypertension. We found that high consumption of red meat was associated with increased odds of hypertension in West African adults. This is consistent with previous meta-analyses involving seven cohort studies with 97 745 hypertension cases, albeit with higher heterogeneity  . Similar) [et al. conducte-N-oxide –95. ThesOur meta-analysis of five cross-sectional studies supports this evidence and indicates that in West Africa, high consumption of dietary fat is associated with increased odds of hypertension. This agrees with prospective cohort studies that have reported increased odds of hypertension (up to 40%) in population subgroups that consume high quantities of dietary fat ,96. AlthOur meta-analysis suggests that higher consumers of dietary salt in West Africa are more likely to be hypertensive compared with limited salt consumers. Our findings are consistent with previous meta-analyses that report an increased risk of hypertension among individuals with a high dietary salt intake that live in rural and urban populations of low to middle-income countries , and an n = 361 254 participants and 90 160 incident cases of hypertension) reported that alcohol consumption increases the risk of hypertension between men  and women  who drank more than two drinks per day in comparison with abstainers with a significant amount of heterogeneity [Finally, our meta-analysis found that high alcohol consumption, exceeding two drinks per day for men and one drink per day for women, is associated with an increased odds of hypertension in West Africa. This finding is consistent with previous studies conducted outside West Africa, which reported an unfavourable association between alcohol consumption and hypertension. A meta-analysis of 20 cohort studies  may have limited our ability to uncover some associations. Nonetheless, our study has several notable strengths worth mentioning: this is the first systematic review and meta-analysis to evaluate the association between dietary factors and hypertension across West Africa. This unique focus enables a more compressive overview of the current state of research in West Africa; despite focusing on a specific region of Africa, a large number of studies and participants were included in this meta-analysis, providing a comprehensive understanding of the topic. This inclusion allows for a broad analysis of the impact of diet on hypertension; a diverse selection of dietary factors was identified that reflected foods groups common to healthy and unhealthy patterns enabling a broad analysis of the impact of diet on hypertension; and analyses were critically appraised for risk of bias and power to ensure high-quality.In conclusion, this systematic review and meta-analysis demonstrated that high consumption of dietary salt, red meat, junk food, dietary fat, and alcohol are associated with an increased likelihood of hypertension in West Africa. In contrast, high fruit and vegetable consumption appears protective. The results of this study will offer regional-specific evidence for developing clinical nutritional assessment tools for clinicians, patients, and researchers and reduce the ever-increasing trend of hypertension in West Africa.The authors sincerely thank the Tertiary Education Trust Fund (TETFund), Nigeria, for funding this study. They also acknowledge the contributions of all the authors and researchers whose work was included in this systematic review and meta-analysis. Finally, they thank all the volunteers in included studies across several countries in West Africa for their dedication and commitment.This study was funded by the Tertiary Education Trust Fund (TETFund) of Nigeria. MAZ is currently funded by Wellcome Trust (217446/Z/19/Z).There are no conflicts of interest.


FSHR polymorphisms in PCOS but the findings are conflicting. A meta-analysis was carried out to decipher the emerging perspectives.Two polymorphisms, rs6165 and rs6166 located in the intracellular domain of FSHR has been reported to affect folliculogenesis, steroidogenesis and oocyte maturation. Several studies have highlighted the role of A comprehensive literature search was made using PubMed, PCOSkb, and Google Scholar. New Ottawa Scale has been utilized to evaluate the quality of each article. To evaluate the strength of association under different genetic models of rs6165 and rs6166 polymorphisms, odds ratio with a 95% confidence interval (CI) was calculated.p = 0.006, OR = 0.65, CI: 0.48–0.89, p = 0.006, OR = 0.82, CI: 0.7–0.95, p = 0.01, respectively) and low to moderate risk in the Caucasian population under allele model .A total of 20 articles were selected for the present study. In pooled analysis and after the stratification by ethnicity, polymorphism rs6165 remains unrelated to the onset of PCOS. Besides, rs6166 exhibits significant protection in the Indian population under recessive, additive, and allele models (OR = 0.7, CI: 0.54–0.9, This meta-analysis suggests that GG genotype of rs6166 provides protection against PCOS, in a population-specific manner. Polycystic ovary syndrome (PCOS) is a complex endocrine disorder affecting females of reproductive age and is the foremost cause of anovulatory infertility , 2. The The precise etiology of PCOS is yet unknown, but it was suggested that the interplay of genetic and environmental factors is responsible for this condition . In PCOSFSHR specifically in exon 10 can lead to the arrest of follicle development at several phases of growth and has several effects on phenotype such as variable development of secondary sex characteristics, primary amenorrhea, hypoplastic ovary and high serum levels of FSH [FSHR and are well known to affect the efficacy of FSHR receptor towards its ligand (FSH), increase FSH levels in a compensatory manner. This increases FSHR resistance leading to reduce estrogen and inhibin B that establish the inhibitory feedback loop in the pituitary gland, resulting in hyperandrogenism which may arrest follicle development [FSHR polymorphisms [The FSHR is located on chromosome 2 at position p21–p16 and has 10 exons and 9 introns. The extracellular domain of the receptor is encoded by the first 9 exons. In contrast, the C-terminal end of the extracellular domain, the whole transmembrane domain, and the intracellular domain of the FSHR are all encoded by exon 10. In females, FSHR is expressed in granulosa cells and regulates the development of graffian follicles, granulosa cell proliferation and estrogen synthesis . When FSs of FSH , 16. Serelopment . Accordielopment , 19. Sevelopment . Althougelopment , howeverorphisms , 25. Furorphisms  includesFSHR polymorphisms (rs6165 and rs6166) without any language barrier (up to March 2023). The following set of MeSH keywords and terms were used: Polycystic Ovary Syndrome or PCOS or Stein Leventhal Syndrome; FSHR or Follicle Stimulating Hormone Receptor or FSH; rs6165 or rs6166 or Ala307Thr or Ser680Asn; gene or allele or genotype or mutation or variant or variation or polymorphism or Genetic variant or Genetic variation. Moreover, manual screening was done on the reference lists of research articles and earlier meta-analyses.Comprehensive computer-based literature searches on Google Scholar, PubMed, and PCOSkb were used to find each study that has reported the genetic association of Polycystic Ovary Syndrome and In this meta-analysis, studies fulfilling the following criteria were included: (a) a case–control design (b) evaluation of the association of rs6165 and rs6166 with PCOS (c) genotype frequency of controls in the Hardy Weinberg equilibrium (HWE) (d) provides genotypic data for both cases and control group (e) studies on human blood samples (f) published in the English language. Exclusion criteria were as follows: (a) not the case–control design, (b) Controls genotype frequency deviated from HWE (c) The design is based on family or sibling pairs (d) Animal studies.Data were extracted from selected publications based on inclusion criteria. From each study, the following information was gathered: first Author name, year of publication, country of origin, diagnostic criteria of PCOS, method of genotyping, the total number of cases and controls and evidence of HWE in controls.To check the quality of each publication included in the present meta-analysis New Castle Ottawa scale (NOS)  was used2 statistic was used and a random effect model (REM) was chosen when I2 was greater than 50%, indicating that heterogeneity is present, while a fixed effect model (FEM) was chosen when I2 was less than 50%, indicating that heterogeneity was absent. To assess publication bias, Beggs’s funnel plot was used. All data were analysed using Review Manager 5.4.1. Bonferroni correction was applied to p-value in order to reduce the type 1 error.The power of the study was calculated using a Cats-power calculator which rendered the power to be > 95%. A goodness of fit Chi-square calculation was used to determine any deviations from HWE. Pooled odds ratios (OR) and 95% confidence intervals (CIs) were used to evaluate the strength of the association for the meta-analysis. The association was determined using the following four genetic models: dominant model (GG + AG vs. AA), recessive model (GG vs. AG + AA), additive model (GG vs. AA), and allele model (G vs. A). To assess heterogeneity, the Ip-value of included studies were tabulated in Table Figure p = 0.49; recessive model: OR-1.19 CI:1.03–1.3, p = 0.02; additive model: OR-1.2 CI:1.02–1.42, p = 0.03, allele model: OR = 1.07, CI: 0.99–1.18, p = 0.08 respectively) ; however, genetic models did not indicate any risk for the development of PCOS  , while dominant model showed no association . Furthermore, a significant association was also found under the allelic model in other studies (Caucasian studies) . However, none of the genetic models show any association with the Asian population  polymorphisms with PCOS risk. It comprises a total of 20 studies (Table p = 0.01), however, in the Asian population, rs6166 polymorphism remained non-associated.A meta-analysis is a statistical tool used to combine the findings of multiple studies on the same topic, increasing the statistical power to resolve discrepancies, and  as well , 22. Addes Table . For rs6FSHR polymorphisms. Another meta-analysis was carried out by Qiu et al. [Chen et al.  conducteu et al.  with 11 u et al.  also didFSHR polymorphisms and PCOS risk. We scored each article using the Newcastle–Ottawa Quality Assessment Scale in order to find higher-quality publications, and each study included in the current meta-analysis received a rating of at least five. We examined all the included studies for fixed or random effect models and analyzed the total effects under dominant, recessive, additive, and allele models. This meta-analysis includes a higher number of studies than the earlier ones, thus it provides accurate estimation. In the present meta-analysis, it was concluded that polymorphism rs6166 was found to have a modest impact on PCOS, however, on a specific cohort. In a meta-analysis, if the degree of heterogeneity rises, it gets harder to justify an integrated conclusion. Heterogeneity for rs6166 polymorphism is higher despite of subgroup analysis, therefore these results cannot be generalized. Further studies on homogeneous and larger populations with ethnicity-matched controls are required to strengthen the statistical power and to better understand the role of FSHR polymorphisms with PCOS.Our meta-analysis is the most comprehensive on the


Observational studies have reported that educational attainment has been related to the risk of esophageal cancer (EC) and its precancerous lesions. However, the causal relationship remains controversial. We aimed to apply the Mendelian randomization (MR) design to determine the causal associations between genetically predicted educational attainment and EC, Barrett's esophagus (BE), and gastroesophageal reflux disease (GERD), and to explore whether modifiable risk factors play a mediating role.n = 293,723), college completion , EC , BE , and GERD , genetic associations between two education phenotypes and EC, BE, and GERD were tested by two-sample MR analyses. Then, two-step MR mediation analyses were used to assess the proportion of the aforementioned association that might be mediated by body mass index (BMI), major depressive disorder (MDD), smoking, drinking, carbohydrates, fat, and protein intake.Using summary statistics from genome-wide association studies (GWASs) based on European ancestry individuals of several years in education , 0.64 [95% confidence interval (CI) 0.44–0.94], 0.86 , and 0.62 }. EduYears was negatively associated with BMI, MDD, and smoking (range of OR: 0.76–0.84). There were positive associations between BMI, smoking with EC, BE, and GERD, as well as between MDD with GERD (range of OR: 1.08–1.50). For individual mediating effect, BMI and smoking mediated 15.75 and 14.15% of the EduYears-EC association and 15.46 and 16.85% of the EduYears-BE association. BMI, MDD, and smoking mediated 5.23, 4.98, and 4.49% of the EduYears-GERD association. For combined mediation, the aforementioned mediators explained 26.62, 28.38, and 11.48% of the effect of EduYears on EC, BE, and GERD. The mediating effects of drinking and dietary composition were not significant in the effect of education on EC, BE, and GERD.Our study supports that genetically predicted higher educational attainment has a protective effect on EC, BE, and GERD, and is partly mediated by reducing adiposity, smoking, and depression. Esophageal cancer (EC) is the eighth most common malignancy and the sixth leading cause of cancer-related deaths worldwide, with an estimated 604,100 new cases and 544,076 deaths in 2020 . The priPrevious observational mediation studies showed that modifiable risk factors partially mediated the association between education and other diseases , 10. EC,Due to the methodological limitations of traditional observational research, such as uncertain temporal relationships, insufficient sample sizes, short follow-up periods, or potential confounding factors, the foregoing associations may not stem from an underlying causal effect. In addition, given the long interval between education and EC, randomized controlled trials (RCTs) are difficult to implement in this field. Therefore, it is necessary to improve causal inference through other study designs. Mendelian randomization (MR) uses randomly assigned genetic variants [single-nucleotide polymorphisms (SNPs)] during conception as instrumental variables for exposure, which could minimize measurement errors, confounding, and reverse causality . The pubIn this study, we aimed to access whether there is a potential causal association between educational attainment and EC, BE, and GERD. In addition, we aimed to access whether modifiable risk factors mediate the effect of educational attainment on EC, BE, and GERD.This study used summary data published by multiple GWAS; ethical approval and patient consent were obtained by corresponding studies –26. ThisWe performed univariable, multivariable, and two-step two-sample MR mediation analyses to investigate whether genetically predicted educational attainment was causally associated with the risk of EC, BE, and GERD, and to assess the proportion mediated by modifiable risk factors in the aforementioned associations. The flowchart of MR analysis in this study is shown in For educational attainment, we used two exposure phenotypes including the number of years in education   and collN = 249,752) is defined as the average number of cigarettes smoked per day by current or former smokers. Drinks per week  is defined as the average number of all types of alcohol consumed per week, which reflected common alcohol drinking behavior, not excessive or harmful drinking behavior. For dietary factors, we utilized the GWAS meta-analysis of European ancestry individuals  ], psychological factor , lifestyle factor , and dietary factor . Genetic association results for BMI were extracted from GWAS meta-analysis published by the Genetic Investigation of ANthropometric Traits (GIANT) Consortium, totaling 322,154 individuals of European ancestry . Genetic268,922) .Summary-level genetic data for EC, BE, and GERD was obtained from the UK Biobank which was a population-based cohort study that recruited more than 500,000 volunteers aged 40–69 years between 2006 and 2010. We used the second-round analysis of UK Biobank data from the Pan-ancestry genetic analysis of the UK Biobank (Pan-UK Biobank). In this project, participants have been divided into six ancestry groups to account for population stratification, and we only restricted the analysis to individuals of European ancestry to avoid bias caused by population stratification. EC , BE , and GERD  with European ancestry were extracted from the Pan-UK Biobank website and IEU OpenGWAS project website . All disThe two-sample MR is an approach that uses a set of SNPs as instrumental variables (IVs) to estimate the causal effect of exposure or mediator on the outcome, where the valid IVs need to satisfy three core assumptions: (1) the relevance assumption, instrumental variables were associated with the exposure, (2) the independence assumption, instrumental variables were not associated with any confounders of the exposure-outcome association, (3) and the exclusion restriction assumption, instrumental variables were only associated with the outcome through the exposure and potential mediators .P < 5e-8) were identified. Independent SNPs loci were obtained by clumping with a threshold of linkage disequilibrium (LD) r2 < 0.001 and a distance of 10,000 kb. Since the explained variance (R2) of almost all IVs to exposure under a strict criterion was <1%, SNPs using a more relaxed statistical threshold (P < 5e-5) were extracted to improve the explained variance of IVs. This relaxed statistical threshold of the genetic IVs approach had been described in previous MR studies  by setting action = 2   . Finally website . If MR p website . The fin website .Q statistic and I2 index were used to test for the presence of heterogeneity  , with q < 0.05 for FDR as significant evidence of associations, and between q > 0.05 for FDR and uncorrected P < 0.05 as suggestive evidence of associations  . If Cochciations . This meciations .I2 and Cochran's Q were used to further assess whether there was heterogeneity. If there was significant heterogeneity, the multivariable QHET method was used; otherwise, the multivariable IVW method was used for MR analysis  of these SNPs of IVs from GWASs of exposure, mediator, and outcome, and harmonized effect allele. Third, a multivariable MR-Egger intercept test was performed to test for potential pleiotropy. If there was significant pleiotropy, a multivariable MR-Egger method was used for MR analysis . When thanalysis .The directed acyclic graph of the MR mediation analysis is shown in P < 5e-5) and improve statistical efficiency and estimation accuracy of models, we applied the Radial IVW method (P < 5e-8). Third, to verify the representativeness of the IVs of EduYears in the main analysis, the MR analysis was repeated using education with the largest GWAS whereas with a large overlap with the outcome sample. Finally, a leave-one-out sensitivity analysis was performed by eliminating a single SNP from the MR analysis one by one to assess the effect of individual variation on causal estimation.A series of sensitivity analyses were performed to ensure the robustness of our findings. First, complementary MR analysis methods , BE , and GERD . Secondary analysis results showed that genetically predicted 1-unit-higher log-odds of liability to college completion were associated with significantly decreased risks of GERD . However, there was no evidence that college completion significantly decreased the risk of EC and BE.P > 0.05). Genetically predicted EduYears was significantly negatively associated with four risk factors, including BMI , MDD , cigarettes smoked per day , and fat intake . Similarly, secondary analysis results showed that genetically predicted College was significantly negatively associated with four risk factors, including BMI , MDD , cigarettes smoked per day , and fat intake .P > 0.05). Genetically predicted BMI significantly or suggestively increased the risks of EC, BE, and GERD . Genetically predicted MDD significantly increased the risks of GERD . Genetically predicted smoking significantly or suggestively increased the risks of EC, BE, and GERD .The UVMR analysis results of the effects of modifiable risk factors on EC, BE, and GERD are shown in The robustness of the main results through a variety of supplementary MR analyses was verified –S13. TheTo the best of our knowledge, this is the first comprehensive MR study to examine the causal effects of education on EC, BE, and GERD, and to access the potential mediational effect of modifiable risk factors on the causal association between education and EC, BE, and GERD. Our results suggested that genetically determined higher levels of education were causally associated with decreased risk of EC, BE, and GERD. In addition, we found that genetically predicted higher levels of education were causally associated with lower BMI and lower risk of depression, as well as the lower frequency of smoking. In the aforementioned causality, BMI and smoking mediated 15.75 and 14.15% of the total effect of EduYears on EC, and 15.46 and 16.85% of the total effect of EduYears on BE. The effect of EduYears on the risk of GERD was mediated by BMI, MDD, and smoking, with the proportion mediated at 5.23, 4.98, and 4.49%. For combined mediation, the aforementioned mediators explained 26.62, 28.38, and 11.48% of the effect of EduYears on EC, BE, and GERD.N = 425,613) showed that college completion does not significantly reduce the risk of EC  and cigarettes per day (n = 4) with low explained variation and statistical power  and overall cancer morbidity and mortality to modifiable risk factors, among them, obesity, smoking, excess alcohol intake, and low fruit-and-vegetable intake accounted for 11.6, 15.5, 7.7, and 7.7% . This suEducational attainment is a proxy for overall SEP because it is associated with both income and occupation later in life . Low eduThere are three main strengths in the present study. First, the use of SNPs as a genetic tool reduces bias due to confounding and reverse causality, thereby providing precise causal estimates of the effect of education on outcomes. For example, observational studies are susceptible to socioeconomic and other environmental factors when exploring the association between education and esophageal diseases. Mendelian randomization with genetic variants as instrumental variables is generally not susceptible to confounding by behavioral, socioeconomic, and physiological factors . Second,P < 5e-8). Fifth, the statistical power for the outcomes may be limited in the present study. On the one hand, to avoid sample overlap bias, EduYears was derived from the GWAS without UK Biobank individuals. However, the representativeness of the IVs of EduYears had been verified in the sensitivity analysis. On the other hand, due to the limitations of the UK Biobank data, the number of esophageal cancer cases is relatively small; thus, further studies of EC with larger sample GWASs will help to confirm the precision of our results. Finally, the current MR method for mediation analysis assumes a linear association between the exposure, mediator, and outcome (However, there are several limitations to this study. First, although educational attainment is often used as a proxy for SEP, they are not interchangeable. SEP encompasses more factors than education. The results of other SEP factors may be different from our results. However, educational attainment is largely determined early in life and comparatively easy to measure, while other SEP factors including occupation and income may be influenced by disability at work or other factors . Second, outcome –72, whic outcome , 73–76. Our MR study supports a protective effect of higher educational attainment on the risk of EC and its precancerous lesion and disease including BE and GERD. Moreover, our study provides genetic evidence that the causal effect of educational attainment on EC, BE, and GERD is partly mediated by BMI, smoking, and MDD. These findings demonstrate that improving metabolic, lifestyle, and psychological factors in low-education populations will have important public health significance for the prevention of EC. However, the majority proportion of education's effect on esophageal cancer and its precancerous lesion and disease remains unexplained. Further research into potentially modifiable mediators is necessary.The original contributions presented in the study are included in the article/XZ and ML conceived and designed the project and interpreted the results. XZ analyzed and cleaned the data and wrote the manuscript. XYa, TZ, XYi, JM, and ML critically reviewed and revised the manuscript. All authors have approved the final manuscript.


Existing evidence on the association between sedentary behavior (SB) and cognitive function remains inconclusive. Therefore, this study investigated the association between SB and the risk of cognitive decline (CD) or mild cognitive impairment (MCI) in the elderly.A comprehensive search was independently conducted by two researchers (XC and GQ) in seven electronic databases, including Medline (via PubMed), China Biology Medicine, Embase, Web of Science, China National Knowledge Infrastructure, Wanfang database, and VIP database for Chinese technical periodicals, covering studies published from the inception of database to June 2023. Studies that investigated the relationship between SB and the risk of CD or MCI in the elderly were included. The quality of the literature was assessed using the Newcastle–Ottawa Scale (NOS) and the Agency for Healthcare Research and Quality (AHRQ) assessment tools. The combined effect size analysis, subgroup analysis, and publication bias assessment were performed using STATA 14.0.A total of 13 cross-sectional and 6 cohort studies involving 81,791 individuals were included, comprising 17 high-quality studies and 2 medium-quality studies. We found that SB was significantly associated with an increased risk of CD  or MCI  among the elderly. Subgroup analysis stratified according to comorbidity, lifestyle, family structure, publication year, and region showed statistical differences between groups, and the consistency of the results revealed the sources of the heterogeneity.This meta-analysis showed that SB is positively associated with the risk of CD or MCI in the elderly, providing a higher level of evidence for the promotion of healthy behaviors by clinicians and health policymakers. Due to the number and quality of the included articles, more high-quality longitudinal studies are needed to further confirm our findings. Cognitive decline (CD) refers to the measurable deterioration observed in various domains of cognitive function, including memory, language, and reasoning  defined SB as all activities with ≤ 1.5 metabolic equivalents (METs) of energy expenditure, such as sitting or lying down  and the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) , China Biology Medicine, Embase, Web of Science, China National Knowledge Infrastructure, Wanfang database, and VIP database for Chinese technical periodicals. They were systematically searched to identify relevant articles published from inception to 11 June 2023. The search was performed using a combination of free words and theme words. Search strategies were developed using Boolean logical operators and truncates. The main search terms included “cognitive dysfunctions,” “mild cognitive impairment,” “cognitive decline,” “cognition,” “sedentary behavior,” “sedentary lifestyle,” “physical inactivity,” and “sedentary time”  principles of evidence-based medicine  extracted data from the included studies separately, and disagreements were resolved by consensus or consultation with a third researcher. Data extracted from the articles included the year of publication, type of study, follow-up period, region, age, sex ratio, sample size, exposure characteristics, and outcome indicators. In addition, when there were several studies in the same article, the data were extracted separately, and in the case of incomplete data, the authors were contacted to request the complete data. Quality assessment of the included studies was conducted using the Newcastle–Ottawa Scale (NOS) and the Agency for Healthcare Research and Quality (AHRQ) , region (developed countries vs. developing or undeveloped countries), sample size , year of publication (≥2017 vs. <2017), length of SB exposure (≥5 vs. <5 h/day), complications (yes vs. no), lifestyle , comorbidity , and family structure . In addition, statistical significance was evaluated by testing for differences across the effect sizes of each subgroup using a 95% CI. Egger's test was used to evaluate publication bias in the included studies, and a P-value of >0.05 hypothetically indicated that there was no publication bias.Meta-analysis was performed using STATA 14.0 software. The combined odds ratios (OR) and 95% confidence intervals (CI) were used to evaluate the relationship between SB and the risk of CD or MCI in the elderly, with a A total of 9,713 relevant articles were retrieved from each database, and 1,699 duplicate articles were removed. After the initial screening of titles and abstracts, 7,782 articles were excluded, and 232 were selected for full-text screening. Furthermore, 213 articles did not meet the criteria and were excluded. Finally, 19 articles  were included ; thus, the fixed effect model was used for meta-analysis. The combined meta-analysis results demonstrated a significant difference between the two groups , and long-term SB was found to increase the risk of CD in the elderly compared to the control group, as shown in Five articles investigated the relationship between SB and the risk of CD in the elderly, with a sample size of 8,439 cases , and a random effects model was used for the meta-analysis. The combined results showed a significant difference between the two groups , and prolonged SB was found to increase the risk of MCI in the elderly compared with control groups, as shown in In total, 14 articles investigated the relationship between SB and the risk of MCI in the elderly, with a sample size of 73,352 cases  compared to that within 5 h/day  though no difference was observed between these two subgroups. Detailed results for other subgroup analyses are shown in No subgroup statistical differences were observed in the relationship between SB and the risk of CD across the two groups or were any significant sources of heterogeneity identified. Significant statistical differences were observed in most subgroups regarding the relationship between SB and the risk of MCI. Furthermore, the consistency among these subgroups provided insights into the sources of heterogeneity. Subgroup analysis of SB exposure time showed significantly higher heterogeneity for sedentary time above 5 h/day  or had short-term SB, and this trend is even more evident in Asian populations.This study showed that prolonged SB increased the risk of CD in the elderly , and there was a significant positive correlation between prolonged SB and increased risk of CD. Related studies have also revealed results consistent with ours , Mendelian randomization (MR) studies, or basic research would be needed.In summary, our study reveals a positive association between SB and the risk of CD or MCI in the elderly. Long-term SB increases the risk of CD or MCI in the elderly. We recommend that the elderly reduce their SB time and increase their level of PA to promote healthy cognitive aging. Considering the quantity and quality of the included articles, our findings need to be interpreted with caution, and more high-quality longitudinal studies are required in the future to further demonstrate the association between SB and the risk of CD or MCI in the elderly.Data generated or analyzed during this study are included in this published article or in the data repositories listed in the references.X-yC served as the principal author, had full access to all data in the study, and took responsibility for the accuracy of the data analysis and the integrity of the data. J-hL and FW contributed to the conception and design. M-yZ, G-pQ, and X-yC contributed to data acquisition and interpretation. X-yC and G-pQ contributed to the draft of the manuscript. Y-jD and J-hL revised the article and finally approved the study. All authors contributed to the article and approved the submitted version.


Both osteoporosis and cognitive impairment affect overall health in elderly individuals. This study aimed to investigate the association between cognitive impairment and the risk of osteoporosis.PubMed, Web of Science, and the Cochrane Library were searched for studies on the association between osteoporosis and cognitive impairment from their inception until August 2023. The random-effects model was used to calculate the pooled risk ratio (RR) of osteoporosis in patients with cognitive impairment. Subgroup analysis was used to detect the sources of heterogeneity. Sensitivity analysis was used to test the robustness of the pooled results. Funnel plots, Egger’s test, and Begg’s test were used to test publication bias.p < 0.001). Subgroup analysis showed that patients with Alzheimer’s disease (AD) are at 1.7-fold risk of osteoporosis compared with the control group , and sex, cognitive classification, study region, study design, and study quality might be the sources of heterogeneity. Sensitivity analysis showed robustness of the pooled results. No significant publication bias was found .Ten studies involving 9,872 patients were included in this meta-analysis. The pooled results showed that patients with cognitive impairment had an increased risk of osteoporosis (RR = 1.56, 95% confidence interval [CI]: 1.30–1.87, Current evidence suggests that patients with cognitive impairment are at increased risk of osteoporosis, especially patients with AD.The online version contains supplementary material available at 10.1186/s12877-023-04548-z. Osteoporosis is the most common metabolic bone disorder and is characterized by a reduction in bone mineral density (BMD), deterioration of bone microarchitecture, and accumulation of marrow fat, which subsequently leads to increased bone fragility and fracture susceptibility , 2. OsteNeurodegenerative disorders, such as Alzheimer’s disease (AD), Parkinson’s disease, and amyotrophic lateral sclerosis, are characterized by the progressive degeneration of neurons in the central nervous system. Numerous studies have extensively investigated the contribution of neurodegenerative disorders in various peripheral disorders –10. MoreAD, the most common neurodegenerative disorder, is mainly characterized by a progressive decline in memory and cognitive function and accounts for 75% of all dementia cases , 24. Glop < 0.01)  OR dementia [Title/Abstract] OR cognitive impairment [Title/Abstract]) AND (osteoporosis [Title/Abstract] OR bone loss [Title/Abstract] OR bone density [Title/Abstract]). The search strategies were adjusted accordingly to suit each database. Furthermore, we also manually checked the reference lists of related articles to identify additional eligible studies.EndNote X9 software was used to manage the literature. We imported literature entries into the software and then removed the duplicate literature. Ineligible studies were excluded by an initial screening of the title and abstract. Finally, we determined eligible studies that met our eligibility criteria based on a full-text review of the remaining studies.The following data were extracted into a standardized Microsoft Excel spreadsheet: first author, year of publication, country of origin, study design, sample size, age, sex, tool of cognitive assessment, site of osteoporosis assessment, and number of patients with bone loss (osteoporosis and/or osteopenia).The methodological quality (MQ) of the selected studies was evaluated using the Newcastle‒Ottawa Scale (NOS) . In thisThe primary outcome was the incidence of osteoporosis (T Score ≤ − 2.5). In addition, we collected data on osteopenia (− 2.5 < T Score ≤ − 1.0) and bone loss (T Score ≤ − 1.0). According to the World Health Organization criteria, in the clinic, a T Score > − 1.0, − 2.5 < T Score ≤ − 1.0, or T Score ≤ − 2.5 was referred to as normal, osteopenia, or osteoporosis, respectively .I2 statistic, and an I2 value ≥ 50% indicated significant and substantial heterogeneity. Given the significant heterogeneity, a meta-analysis was performed using a random-effects model. Second, the risk ratio (RR) and 95% confidence interval (CI) were determined to calculate the effect sizes. Third, subgroup analysis and sensitivity analysis were used to explore sources of heterogeneity. Finally, a funnel plot was used to detect publication bias [p < 0.05.The meta-analysis was conducted in the following steps. First, study heterogeneity was measured using the ion bias . In addiA total of 5648 studies were retrieved, including 1303 studies in PubMed, 4212 studies in Web of Science, and 133 studies in the Cochrane Library. After 1023 duplicate studies were removed, we read the titles and abstracts of the remaining 4625 studies and excluded studies that did not meet our eligibility criteria. We then reviewed the full text of the 41 studies and further excluded ineligible studies. Finally, 10 studies that fulfilled the predefined eligibility criteria were included in this meta-analysis. Following the PRISMA 2020 flow diagram, we showed the study screening process in Fig. The basic characteristics of these 10 studies are summarized in Table The results of MQ evaluation based on the NOS tool are shown in Table I2 = 88.0%, p < 0.001); thus, we used a random-effects model to perform the meta-analysis. The pooled results showed that patients with cognitive impairment were at higher risk of osteoporosis than the control group .Ten studies with a total sample size of 9,872 reported data on the incidence of osteoporosis. The overall prevalence of osteoporosis in patients with cognitive impairment was 42.8%, while in the control group, it was 27.6%. As shown in Fig. I2 = 58.2%, p = 0.035); the pooled results from a random-effects model showed that patients with cognitive impairment had a higher risk of osteopenia . As shown in Fig. I2 = 75.3%, p = 0.001), we used a random-effects model to pool the results and found that patients with cognitive impairment had a higher risk of bone loss .Six studies provided data on the incidence of osteopenia and/or bone loss. As shown in Fig. I2 values significantly changed in the subgroups of sex, cognitive classification, study region, study design, and study quality, which indicated that these factors might be the sources of heterogeneity. The results of the subgroup analysis are shown in Table p = 0.001).Subgroup analysis of the risk of osteoporosis was performed according to different factors, such as age, sex, cognitive classification, region, sample size, study design, and study quality. We found that I2 = 88.0%, p < 0.001), a sensitivity analysis based on the leave-one-out strategy was performed to test the robustness of the pooled results of the risk of osteoporosis. As shown in Fig. Given that significant heterogeneity existed .Interestingly, brain-derived EVs are also involved in the regulation of bone metabolism. Recently, Liu et al.  found thp < 0.001). However, no significant difference was observed in the other subgroup (p = 0.327). These findings suggest that the association between osteoporosis and cognitive impairment may be influenced by ethnic differences.The subgroup analysis found lower heterogeneity in the cohort study group and the high MQ group, suggesting that more well-designed studies are needed to confirm our findings. We found no significant association between osteoporosis and cognitive impairment in the male subgroup. Since the average age of the included population was > 62 years, loss of estrogen protection might be the major cause of osteoporosis in postmenopausal women. In addition, our subgroup analysis showed that the RR of osteoporosis was 1.40, 1.55, and 1.95 in subgroups ≥ 65 years old, ≥ 60 years old, and > 50 years old, respectively. There are many possible confounders in elderly individuals, which may impact our results on the correlation between osteoporosis and cognitive impairment. Therefore, future research should collect more younger individuals, especially those between 40 and 60 years old. Importantly, among the ten included studies, eight were conducted in Asian countries. Subgroup analysis indicated significant differences in all subgroups of Asian countries (all Our study has several strengths. First, this is the first meta-analysis to support the high risk of osteoporosis in patients with cognitive impairment, especially AD patients. Second, each study included in this meta-analysis had moderate or high MQ, and its sample size was relatively large. Third, our subgroup analysis is multiperspective, and its results may provide novel ideas for future research design. In addition, the results of sensitivity analysis and publication bias testing confirm that our conclusions are robust. Importantly, our results are consistent with previous clinical and basic studies, which contribute to the great confidence of the findings. However, several limitations should be treated with considerable caution. First, the bias caused by different design of study can affect our conclusions. Second, there was considerable heterogeneity among the studies, warranting a careful interpretation of the findings. Third, we did not limit the criteria of cognitive impairment for the included studies. Fourth, all subjects in the current study were over 50 years old, and future research needs to reanalyze the difference in younger individuals. Finally, most of these included studies were conducted in Asian countries, which may weaken the applicability of our conclusions. Current evidence only supports the association between cognitive impairment and an increased incidence of osteoporosis within the Asian population. Although we performed subgroup analysis to clarify the state of the evidence, more well-designed large-scale studies are still needed to validate our findings in the future.In conclusion, our findings support that patients with cognitive impairment are at increased risk of osteoporosis, especially patients with AD. We encourage early treatment of neurodegenerative diseases to improve cognitive function, avoiding bone loss caused by brain aging. However, considering the high heterogeneity between the included studies, more meticulously designed studies are needed.Below is the link to the electronic supplementary material.Supplementary Material 1: Subgroup analyses and publication bias test


Lipid accumulation product (LAP) is an index calculated by waist circumference (WC) and triglyceride (TG), which reflects lipid toxicity. This study aims to investigate the association between the LAP index and nonalcoholic fatty liver disease (NAFLD) in a systematic review and meta-analysis.P < 0.001) of the LAP index. The DTA meta-analysis was conducted and showed that the LAP index pooled sensitivity and specificity for screening of NAFLD were 94%  and 85% , respectively.PubMed, Scopus, and Web of Science online databases were searched for eligible studies that investigated the association of the LAP index and NAFLD. Sixteen observational studies with 96,101 participants, including four cohort studies, one case‒control study and 11 cross-sectional studies with baseline data, were entered into this analysis. Fourteen studies reported a significant association between the LAP index and NAFLD, and two reported that this relation was not significant; two different meta-analyses (1- mean difference (MD) and 2- bivariate diagnostic test accuracy [DTA]) were conducted using Stata version 14. The LAP index was compared in subjects with and without NAFLD, and the difference was significant with 34.90 units (CI 95: 30.59–39.31, The LAP Index is an inexpensive, sensitive, and specific method to evaluate NAFLD and may be valuable for NAFLD screening.The online version contains supplementary material available at 10.1186/s12944-023-01802-6. NAFLD is the leading cause of chronic liver disease around the globe . NAFLD aNAFLD is defined as fat accumulation in 5% of hepatocytes or more, without daily alcohol consumption of greater than 20 g for females and 30 g for males or other causes of fatty liver . NAFLD hBiopsy of the liver is the method of choice for diagnosing NAFLD and liver cirrhosis. However, this method is invasive and is not a suitable tool for follow-up . Other dTreatment of NAFLD is mainly based on lifestyle modifications such as weight loss, a low-calorie diet, and aerobic exercise; these modifications lower hepatic fat accumulation and improve liver metabolism –19. TherLAP is an index for estimation of excessive lipid accumulation, which is calculated by WC and fasting plasma TG: (LAP = (WC (cm) – 65) x TG (mmol/L)) for men, and (LAP = (WC (cm) – 58) x TG (mmol/L)) for women . RecentlAs a low-cost and safe method, the LAP index might be an option for detecting NAFLD compared to invasive and expensive procedures such as liver biopsy and MRI-PDFF. Based on our search, no systematic review regarding the relationship between NAFLD and the LAP index has been performed previously. This systematic review aimed to investigate the articles and assess whether the LAP index is reliable for NAFLD screening.The Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement 2020 guidelines were used to conduct this systematic review .This systematic review protocol was registered in the International Prospective Register of Systematic Reviews (PROSPERO); CRD42022334204.Three online databases, including Scopus, Web of Science, and PubMed, were searched systematically until September 2022. Search strings were relevant to the LAP index and NAFLD .This systematic search was performed without any language limitations.We included human studies based on the following PICO. (a) Population: Adult participants (≥ 18 years old) with NAFLD. (b) Intervention: LAP index. (c) Control: Adult participants without any types of NAFLD. (d) Outcomes: The prognostic performance of the LAP index for the diagnosis of NAFLD.This study included original, observational, and peer-reviewed papers with the mentioned PICO. Additionally, studies reporting at least the means of the LAP index among subjects with and without NAFLD were eligible. Additionally, articles detecting NAFLD only by a common diagnostic method  were included.In this study, articles with inadequate information, such as sample size  or standard deviation (SD), were not eligible for this study. In addition, studies with a specific disease or condition as inclusion criteria  were not eligible for this meta-analysis since it was performed on the general population. Additionally, review articles, conferences, preprint papers, abstracts, dissertations, reports, randomized control trials, editorials letters, and chapters were not included.Based on the aforementioned criteria, two independent authors (M. S and M. E) screened the title/abstract, and then, for the remaining records, full texts were reviewed. Regarding the dispute, the two reviewers discussed reaching an agreement. If they could not reach an agreement, the third reviewer resolved disagreements (SA. N).Two independent reviewers extracted data (M. E and R. AB) from each included publication. If they could not reach an agreement regarding an item, the third reviewer resolved disagreements (SA. S).The following data were extracted: the first author’s name of the publication, publication year, design of the study, location of study, important exclusion and inclusion criteria for each study, total case and control number, LAP index mean, and SD of case and control groups, the female to male ratio, and the range age and mean age of subjects. Additionally, the specificity and sensitivity of the LAP index for diagnosing NAFLD were written in the predesigned Google worksheet. In addition, if the LAP index was reported in mg/dl units, it was converted to mmol/l for meta-analysis.The Newcastle–Ottawa Scale (NOS) was used to evaluate the quality of the included studies. This tool evaluates three main elements, including 1- sample selection, 2- comparability of the sample based on the analysis or design, and 3- how the exposure was defined or how outcomes of interest were diagnosed . StudiesStatistical software for data science version 14 through the "Midas", "Metandi" and "Metan" commands were utilized for DTA and MD meta-analysis. The analysis was performed via a random-effects model (a model considered that the true effect might be different from one study to another due to the heterogeneity and differences among studies). Studies that computed the LAP index for the diagnosis of NAFLD were recorded in the MD meta-analysis. If the SDs were not declared in a study, the interquartile range (IQR) or 95% confidence interval (CI) was used to compute the SD using the formulas listed below :2 index were applied to assess the heterogeneity of the meta-analysis. Significant heterogeneity of data was defined as I2 > 50% or a significant Cochran-Q test (P < 0.10). To evaluate the effects of possible confounding factors on the heterogeneity among publications, subgroup analyses were performed. NAFLD diagnostic method and geographic area (country and continent) of studies were used for subgroup analysis as categorical variables. Additionally, meta-regressions on continuous variables  were performed.Studies that calculated the specificity and sensitivity of the LAP index for diagnosing NAFLD were selected for DTA meta-analysis. Bivariate DTA meta-analysis was conducted to calculate pooled sensitivity and specificity; this model is applied when different cutoffs are reported in studies. The Cochran-Q test and IP < 0.05 [Publication bias was assessed through funnel plots and Begg’s, Egger’s and Deek’s tests with significance levels at P < 0.05 , 32.After the primary search and eliminating duplicate records, 122 unique results were identified. In the next step, 39 studies were excluded by screening titles or abstracts. Throughout screening by titles/abstracts, reviewers had an extremely conservative approach. Finally, 83 remaining studies underwent a full-text evaluation. Based on the remaining articles' relevance to this study's purpose and the predefined eligibility criteria for this review, 16 studies were selected for this study , and only one study used Chinese diagnostic criteria to detect patients with NAFLD [Different methods were used to diagnose NAFLD in studies; the most common one was ultrasonography  showed high heterogeneity in this meta-analysis.The MD meta-analysis (random-effect model) of 16 selected articles demonstrated that the pooled mean LAP index in subjects with NAFLD was 34.90 units (CI 95: 30.59–39.31) higher than that in participants without NAFLD Fig. . The resSubgroup analyses were conducted based on the NAFLD diagnostic method and geographic area of studies Table . AdditioBegg’s and Egger’s regression tests  and visuP = 0.624). Therefore, a bivariate DTA meta-analysis was conducted to evaluate the combined specificity and sensitivity of the LAP index in screening participants for NAFLD.Five studies with 17,934 participants were selected in the DTA meta-analysis , 38, 39.P < 0.001) and specificity at 85%  were calculated.  curve was plotted, and the Area Under the Curve (AUC) was 0.95 (CI 95: 0.93—0.97), which showed that the LAP index is accurate for the screening of NAFLD  sensitivity and 73%  specificity of the LAP index for the screening of NAFLD, and the heterogeneity was reduced significantly based on HSROC. The new sROC curve was graphed with an AUC of 0.82 (CI 95: 0.79—0.85). Based on the pooled odds ratio, the population with a higher LAP index was at 15.52 (10.96 – 21.97) times more risk of NAFLD.The heterogeneity of the sROC curve is defined by visual inspection. Observation of the area between the confidence and prediction regions showed that the heterogeneity of the sROC curve is moderate to high . Based oP = 0.25).Deeks’ funnel plot did not reveal any substantial publication bias regarding the DTA meta-analysis ; however, high heterogeneity was observed in the results. Therefore, we performed subgroup and meta-regression analyses. Subgroup analysis based on geographical areas (continent where the study was performed) reduced heterogeneity  were selected for the DTA meta-analysis to calculate the pooled specificity, sensitivity, and AUC of the LAP index for diagnosing NAFLD. This analysis showed that despite the straightforward calculation of the LAP index, the high specificity and sensitivity of this index for diagnosing NAFLD and candidates it a valuable tool for its screening in populations.This study has some strengths. First, it has extensive and replicable methods for searching published literature. Second, it is the first systematic review and meta-analysis on the LAP index and NALFD in the literature. Third, by including 96,101 individuals with and without NAFLD, a conclusive result with high precision and low bias may have been achieved for the general population in this study. Finally, in addition to resolving the conflict between the existing studies on the relationship between the LAP index and NAFLD, this article introduces the LAP index as an appropriate and low-cost method for NAFLD screening with high sensitivity and specificity.There are some limitations to this study, similar to other systematic reviews. Laboratory data reported in each study were measured in different lab centers with various facilities, and different alcohol consumption limits were defined for each specific study, which may lead to some inconsistency. In addition, suggesting different cutoffs by each study could cause some discrepancies in the final result. Additionally, ultrasonography, an operator-dependent method for diagnosing NAFLD, could result in disparities in interpretation and some cases being missed . Some otThese data support the use of the LAP index for the diagnosis of NAFLD in the general population as an available, low-cost, and accurate tool. Thus, screening and diagnosis of NAFLD can occur more rapidly due to these significant findings in clinical care.Additional file 1: Supplementary Table 1. Newcastle-Ottawa scale stars by domain. Supplemental Figure 1. Forest plot of Lipid accumulation product index mean difference grouped by continent in which study was conducted. Supplemental Figure 2. Forest plot of Lipid accumulation product index mean difference sub grouped by country in which study was conducted. Supplemental Figure 3. Forest plot of Lipid accumulation product index mean difference , sub grouped by diagnostic method of NAFLD. Supplemental Figure 3. Forest plot of Lipid accumulation product index mean difference, sub grouped by design of the study. CS: Cross-sectional. Supplemental Figure 4. Meta regression on Year of publication. Supplemental Figure 5. Meta-regression on female to male proportion. Supplemental Figure 6. Meta-regression on mean age of participants.Additional file 2. Sensitivity Analysis.


Oral cancer is now a top priority for non‐communicable illnesses and universal health care plans, according to the WHO. There is no general estimate of the incidence of oral cavity cancer in Iran, despite multiple investigations. The purpose of this study is to evaluate the age‐standardized incidence rate (ASR) of oral cavity cancers in Iran.Q tests, along with random‐effect models. It was determined what caused the heterogeneity using a meta‐regression model. By eliminating experiments one at a time, sensitivity analysis was used. The meta‐analysis was corrected utilizing the Trim‐and‐fill method due to the identification of noteworthy publication bias via the Egger's test and asymmetry of the funnel plot.In accordance with the MOOSE  Checklist recommendations, this systematic review was conducted. PubMed/MEDLINE, Web of Science, ScienceDirect, Embase, Scopus, ProQuest, and Google Scholar were used as the international databases for the systematic literature search, while SID (Scientific Information Database), Magiran and element were used as the Iranian databases. The heterogeneity of the research will be evaluated by means of the inverse variance and Cochran Q statistic = 1118.09, df = 25, p < .0001, I2 = 97.8%), and 1.46 (95% CI: 1.14–1.77) , respectively. According to the funnel plots and Egger's test, there is no evidence of publication bias in studies reporting on males , but for ASR in females, Egger's test was significant . Based on Trim‐and‐fill methods, overall ASR corrected in females was estimated to be 1.36 (95% CI: 1.05%–1.66%).This research incorporated a total of 22 journal articles. The pooled ASR of oral cavity cancer for males and females was estimated at 1.96 (95% CI: 1.65–2.26) (Iran's oral cavity cancer incidence was lower than the global average, but owing to variables including an aging population, a rise in life expectancy, and exposure to risk factors like smoking, we anticipate an increasing trend. Globally, it was predicted that 10.3 million people will die from cancer in 2020 and 19.3 million people would get the disease. Oral malignancies account for around 3.2% of all cancer‐related fatalities.The primary risk factors for oral cavity cancer include age, male gender, tobacco and alcohol consumption, exposure to UV radiation, immunosuppression (such as from HIV infection or organ transplantation), and human papillomavirus (HPV) infection.The availability of a system for tracking and giving precise information on the incidence of different cancers is one of the crucial elements of the cancer control program. The establishment of the Iran Cancer Registry system dates back to 1984, with the aim of creating a cancer registry at the national level, which would collect and analyze information from various hospitals and health facilities across the country.For more efficient planning and management of cancer prevention initiatives, it is vital to be informed of the existing status and trends in cancer incidence. There have been some comprehensive investigations of the incidence of different malignancies in Iran. As far as we know, there has not been a systematic review or meta‐analysis conducted in Iran to evaluate the incidence and trend of oral cavity malignancies. Therefore, the objective of this study is to conduct a thorough review and meta‐analysis of the age‐standardized incidence rate of oral cavity malignancies in Iran.2https://www.crd.york.ac.uk/prospero/display_record.php?ID=CRD42022374132.As advised by the MOOSE  Checklist,2.1The researchers searched for relevant articles without a time limit till December 2022 in Iranian databases SID (Scientific Information Database), Magiran and element as well as international databases PubMed/MEDLINE, Web of Science, ScienceDirect, Embase, Scopus, ProQuest, and Google Scholar. Additionally, the reference list of the discovered papers was manually reviewed to boost the search's sensitivity. Using ICD‐0‐3 codes, the team determined that the oral cavity consists of the lip , tongue, mouth floor, gum, hard palate, mouth, buccal mucosa, and vestibule. The search method was performed using MeSH terms in combination or separately using “AND,” and “OR” functions (Supplementary Table 2.2The review included studies that provided clear descriptions of the age‐standardized rate (ASR) of oral cavity cancer and contained reports from populations in Iran. Studies with limited sample numbers and those that reported prevalence were excluded. The final meta‐analysis did not take into account review studies or publications presented as conference posters.2.3The reviewers were aware of the name of the publication and the authors to gather information from relevant papers. The age‐standardized incidence rate (ASR), which was taken from the publications and put into this form, the sample size, sex, and the time and place of the research were all included in the data collecting form, which was an Excel sheet that was constructed based on the study's goals. The authors were contacted for access to additional data, and the evaluation of the articles for which the full text was not available.All articles were assessed for duplicates. They were then chosen based on the titles and abstracts of linked papers. To evaluate the quality of the studies, two impartial reviewers assessed the entire text based on the inclusion and exclusion criteria. The cases of disagreement between two referees were independently evaluated, and resolved by the third referee Table .2.4The risk of bias  of the included studies was evaluated using the modified Newcastle Ottawa scale for observational study checklist,2.5I2 statistic and Cochran's Q test. When no heterogeneity was detected, a fixed‐effect model was used. However, in cases of heterogeneity, a random‐effect model was utilized.The meta‐analysis was performed using Stata version 16 software. The researchers assessed the heterogeneity using statistical techniques, such as the The sensitivity analysis approach, which involved one‐by‐one elimination, was used to investigate the influence of each study on the pooled incidence estimate. The researchers assessed the robustness of each model and ultimately selected the best mode.The forest plot was used to display the overall age‐standardized rate (ASR) and its corresponding 95% confidence intervals (CI). To assess publication bias, Egger's and Begg's tests were performed. Additionally, the trim‐and‐fill approach was employed to determine the number of missing studies and adjust the total estimate.33.1n = 24 from Magiran, n = 22 from SID, and n = 14 from Elmnet) and seven international databases  combined to identify 348 studies. The next step. The titles and abstracts of 120 articles and 174 duplicate articles that failed to fulfill the inclusion criteria were removed. Moreover, 32 papers were disqualified from meeting the inclusion criteria, encompassing reviews, letters, case reports, prevalence of other cancers, and accounts of oral cancer prevalence in the complete text. Figure Three local databases , Fars between 1998 and 2002 (0.1 per 100 000), and Ardabil between 2004 and 2006 (0.2 per 100 000) , with random effect model after sensitivity analysis using one‐by‐one elimination of studies and the best robustness model was chosen among the males    Table .3.5Q statistic = 2576.99, df = 26, p < .0001, I2 = 99.0%)  of oral cavity cancer for females, which was determined to be 1.46 (95% CI: 1.14–1.77)  Figure .Q statistic and I2 = 99%. To identify the source of heterogeneity, univariate and multivariable meta‐regression analyses were employed.The high heterogeneity among the studies was observed based on the significant p = .044) and the study period (p = .042) were potentially responsible for the observed heterogeneity among the studies. A multivariable meta‐regression also suggested that the study period may have contributed to the data's heterogeneity . As a result, altering the study period can increase the incidence by 0.04517. In a multivariable analysis, the location, sample size, or bias risk of the studies did not show a significant difference between them .Through a univariate meta‐regression analysis, it was found that sample size studies   Figure .p = .008)  Figure .Based on Trim‐and‐fill analysis using non‐parametric methods, the anticipated values of three censored studies were calculated, and overall ASR corrected by the random effects model was calculated to be 1.36 (95% CI: 1.05%–1.66%) in females.4In Iran, cancer ranks third as a cause of mortality.The findings of this study revealed that the age‐standardized rate (ASR) for oral cavity cancer is 1.96 and 1.36 per 100 000 for Iranian men and women, respectively. According to GLOBOCAN 2020, lip and oral cavity cancer had incidence rates of 6 and 2.3 per 100 000, respectively, on a global scale. Melanesia had the highest ASR , followed by south‐central Asia . Central America, western Africa, and northern Africa have the lowest ASR: 1–1.7 for men and 1.1–1.3 for females. Iran is situated in western Asia, and the ASR for the nations in this area indicated male mortality rates of 2.2 and female mortality rates of 1.3. The findings of this research were near the ASR range for western Asian nations, notably for the female group, which was 1.36. Oral cancer stands among 6th most prevalent malignancies in the world.Similar to the majority of other reports, the ASR ratio for oral cancer showed dominance in men. The reason for the higher incidence of oral cancer in men compared to women could be due to their increased exposure to risk factors, including smoking, alcohol consumption, tobacco use, and outdoor activities that increase exposure to solar radiation. Our result showed a 1.2 age‐standardized incidence rate (ASR) ratio for oral cancer of men to women, however, various numbers are reported in other studies for instance in Taiwan men were predominantly 15 times more than women.In some particular high‐risk countries in South Asia like IndonesiaIn both genders, East‐Azerbaijan province stands among the top 2 with the highest ASR rate. Moreover, Ardebil as the neighboring province to East‐Azerbaijan, both located in the northwest of Iran, has the highest ASR rate for males. Considering both provinces count as high‐risk regions for gastrointestinal cancers, particularly esophageal and gastric cancers,According to the reports, the age‐standardized incidence rate (ASR) of all cancers has been found to be higher in the northern countries of the West Asia region, including the north of Iran, when compared to the southern countries of the region, which includes the south of Iran.Unlike most of the worldOur study had several limitations. First, Although the study shows that the ASR incidence rate for oral cavity cancer in Iran is lower than in the world, taking into the fact that it is reported that all cancers in Iran are underestimated,Increasing awareness to avoid the risk factors exposure, and the development of screening programs will reduce the burden of this disease and its incidence rate.5In Iran, the incidence of oral cavity cancer was lower than the world average. However, we expect an upward trend, as a result of an upward trend in life expectancy, the number of elderly, and various exposures to risk factors, such as smoking. To detect cancer at an early stage, a cancer registry system, and screening programs should be developed.Mohammad Jokar: Data curation ; investigation ; methodology ; software ; visualization ; writing – original draft ; writing – review and editing . Negin Namavari: Data curation ; methodology ; validation ; writing – original draft . Seyed Alireza Moshiri: Data curation ; investigation ; writing – original draft . Hossein Kargar Jahromi: Data curation ; investigation ; visualization . vahid rahmanian: Conceptualization ; formal analysis ; investigation ; methodology ; project administration ; software ; supervision ; validation ; writing – original draft ; writing – review and editing .The authors have stated explicitly that there are no conflicts of interest in connection with this article.In this study, we adhered to all ethical principles in the systematic review and meta‐analysis studies. The Research Ethics Committee of Jahrom University of Medical Sciences approved the study protocol (ID IR.JUMS.REC.1401.112.).The lead author (Vahid Rahmanian) affirms that the manuscript presents a comprehensive, truthful, and clear account of the study. The study plan was adhered to, and any deviations were elucidated, and if necessary, documented.Supplementary table 1 Strategy search for the incidence of oral cavity cancer in Iran.Click here for additional data file.