Secondary constipation refers to constipation that occurs after certain diseases or medications, such as acute stroke or opioids, and the efficacy of electroacupuncture for secondary constipation is controversial. So, this study aimed to explore the efficacy and safety of electroacupuncture for secondary constipation through a meta-analysis and systematic review.We retrieved articles from PubMed, Embase, Cochrane Library, Web of Science, CNKI, Wanfang, and VIP databases up to 28 February 2023. The study was screened strictly according to inclusion and exclusion criteria. Revman5.4 was used for quality evaluation; grade rating was used for index evaluation, and stata15.0 was used for data consolidation analysis.P < 0.001), reduced defecation straining score , increased weekly complete spontaneous bowel movements , and increased in the weekly spontaneous bowel movements , P < 0.001), and electroacupuncture had no effect on change stool consistency score compared , P = 0.88) and did not increase adverse events .Thirteen randomized controlled studies, involving a total of 1437 people (722 electroacupuncture and 715 control groups), were included in this review. Meta-analysis results indicated that electroacupuncture significantly improved constipation overall response (RR = 1.31, 95%CI: 1.11, 1.55, According to the current studies, the overall relief rate of patients with secondary constipation after electroacupuncture treatment was improved, the defecation pressure score was reduced, the weekly natural defecation was more complete, and adverse reactions were not increased. Electroacupuncture therefore shows potential for treating constipation, but more high-quality studies are needed to confirm these findings.The online version contains supplementary material available at 10.1007/s00384-023-04487-6. Constipation is a widespread gastrointestinal disorder, affecting individuals worldwide . With thWestern medicine currently employs gastrointestinal motion-promoting, cathartic drugs, and enema as symptomatic treatment . LactuloThe protocol has been registered in the International Prospective Register of Systematic Reviews database (PROSPERO: CRD42023403707).A systematic literature search was conducted to identify randomized controlled trials on the use of electroacupuncture for the treatment of constipation. PubMed, Embase, Cochrane Library, Web of Science, CNKI, Wanfang, and VIP databases were searched for articles published as of 28 February 2023. The search terms used were “electroacupuncture” and “constipation.” Supplementary materials This study included adults who met the diagnostic criteria for secondary constipation   and utilTwo independent evaluators conducted literature screening for data extraction. The screening process involved reading the title and abstract, as well as the full text of the literature, for those that were easy to judge. For literature that required further evaluation, the full text was downloaded and reviewed. The inclusion and exclusion criteria were strictly followed during the screening process, with the observation indicators of the two groups of studies extracted and cross-checked for consistency. Data extraction included the first author, year of publication, registration number, primary disease, sample size, and follow-up.Two researchers independently assessed the quality of the studies included in the analysis. the Cochrane to Randomized Clinical Trials Risk of Bias Tool 2.0 (RoB2)  was usedGRADE  system  was usedI2 value or Q statistic. I2 values of 0%, 25%, 50%, and 75% indicated no heterogeneity, low heterogeneity, medium heterogeneity, and high heterogeneity, respectively. When I2 was ≥ 50%, sensitivity analysis was performed to explore potential sources of heterogeneity. In cases where heterogeneity was less than 50%, the fixed effects model was utilized. Additionally, the funnel plot was utilized to assess publication bias.The data extracted from the studies was inputted into statistical software, Revman5.4 and Stata15.0 , for analysis. Heterogeneity was tested using either the A total of 604 articles were preliminarily retrieved, 464 articles were acquired after removing duplicates, 32 articles were obtained by checking the titles and abstracts of the articles, and 13 articles –34 were A total of 13 randomized controlled studies were included, involving a total of 1437 people, including 722 electroacupuncture (E) and 715 control groups (C). The specific features of the paper are shown in Table The 13 articles included in this study all explained the method of randomization and the blind method used, and 6 articles , 28–30 eI2 = 77%, P < 0.001), random effects model was used for data analysis. The analysis results , P < 0.001) suggested that electroacupuncture could improve the overall response compared with the control group, Fig. I2 > 50%, sensitivity analysis was conducted on the indicator, and the analysis result indicated that the sensitivity was small and the result was relatively stable, Fig. The overall response is defined as follows: Cure  + improvement  was mentioned in 10 studies, including 535 in the electroacupuncture group and 533 in the control group. A heterogeneity test , random effects model was used for data analysis. The analysis results , P < 0.001) suggested that electroacupuncture could reduce the constipation defecation straining score compared with the control group, Fig. A change in defecation straining score was mentioned in 4 studies, including 170 in the electroacupuncture group and 167 in the control group. A heterogeneity test , random effects model was used for data analysis. The analysis results , P = 0.001) suggested that the electroacupuncture group had a significant increment in the weekly CSBMs compared with the control group, Fig. Changes in the weekly CSBMs were mentioned in 2 studies, including 87 in the electroacupuncture group and 86 in the control group. A heterogeneity test , random effects model was used for data analysis. The analysis results , P = 0.05) suggested that the electroacupuncture had no effect on the CCS score compared with the control group, Fig. I2 > 50%, sensitivity analysis was conducted on the indicator, and the analysis result indicated that the sensitivity was small, and the result was relatively stable, Fig. Change in CCS score was mentioned in 4 studies, including 200 in the electroacupuncture group and 200 in the control group. A heterogeneity test , random effects model was used for data analysis. The analysis results , P = 0.44) suggested that the electroacupuncture had no effect on PAC-QOL total score compared with the control group, Fig. Change in PAC-QOL total score was mentioned in 2 studies, including 133 in the electroacupuncture group and 133 in the control group. A heterogeneity test , random effects model was used for data analysis. The analysis results , P = 0.20) suggested that compared with the control group, there was no increase in adverse event in the electroacupuncture group, Fig. I2 > 50%, sensitivity analysis was conducted on the indicator, and the analysis result indicated that the sensitivity was small, and the result was relatively stable, Fig. An adverse event was mentioned in 8 studies, including 537 in the electroacupuncture group and 531 in the control group. A heterogeneity test  was performed by random effects model. The results of meta-analysis , P = 0.88) suggested that electroacupuncture had no effect on change in stool consistency score compared with the control group, Fig. A change in stool consistency score was mentioned in a total of 4 articles, including 170 samples in the electroacupuncture group and 167 in the control group. The heterogeneity test  was performed by random effects model. The results of the meta-analysis , P < 0.001) suggested that the electroacupuncture could improve changes in the weekly SBMs compared with the control group, and the difference was statistically significant, Fig. Changes in the weekly SBMs was mentioned in a total of 3 articles, including 200 samples in the experimental group and 196 samples in the control group. Heterogeneity test Below is the link to the electronic supplementary material.


Background and objectives: This study aims to elucidate the prognostic implications of Epstein–Barr virus (EBV) infection in gastric carcinomas (GCs) through a systematic review and meta-analysis. Materials and Methods: In total, 57 eligible studies and 22,943 patients were included in this meta-analysis. We compared the prognoses of EBV-infected and non-infected GC patients. The subgroup analysis was performed based on the study location, molecular classification, and Lauren’s classification. This study was checked according to the PRISMA 2020. The meta-analysis was performed using the Comprehensive Meta-Analysis software package. Results: EBV infection was found in 10.4%  0.082–0.131) of GC patients. The EBV-infected GC patients had a better overall survival compared with the EBV-non-infected GC patients (hazard ratio (HR) 0.890, 95% CI 0.816–0.970). In the subgroup analysis based on molecular classification, no significant differences were found between EBV+ and microsatellite instability and microsatellite stable (MSS)/EBV− subgroups . In the diffuse type of Lauren’s classification, EBV-infected GCs have a better prognosis compared with the EBV-non-infected GCs . The prognostic impact of EBV infection was found in the Asian and American subgroups but not in the European subgroup . Conclusions: EBV infection is a favorable survival factor for GCs. However, the prognostic implications of EBV infection in the new molecular classification are not clear. Helicobacter pylori and Epstein–Barr virus (EBV). Burke et al. first reported a correlation between EBV infection and GC in 1990 [Gastric carcinoma (GC) is a common malignancy of the gastrointestinal tract and is a heterogeneous disease with different phenotypes, genotypes, and tumor behaviors ,2. GC ca in 1990 , but the in 1990 ,5,6,7,8. in 1990 ,57,58,59 in 1990 . In addi in 1990 . The EBV in 1990 . Thus, tPublished articles were searched in the PubMed database on 15 August 2022. The keywords used in the search were “gastric carcinoma or gastric cancer or stomach cancer” and “Epstein–Barr virus or EBV”. The titles and abstracts of all searched articles were screened for the inclusion and exclusion of each article. Articles that studied the correlation between EBV infection and the overall survival of GCs were included. Case reports, non-original articles, or articles not written in English were excluded. This study was performed by the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA). The PRISMA checklist is shown in the p-value, and the O–E statistic (difference between the number of observed and expected events) or its variance. If those data were unavailable, HR was estimated using the total number of events, the number of patients at risk in each group, and the log-rank statistic or its p-value. Finally, if the only useful data were in the form of graphical representations of survival distributions, the survival rates were extracted at specified times to reconstruct the HR estimate and its variance under the assumption that patients were followed up at constant intervals [All data were obtained from eligible studies ,57,58,59ntervals . The pubntervals .2 statistics, expressed as p-values. Additionally, we conducted a sensitivity analysis to assess the heterogeneity of the eligible studies and the impact of each study on the combined effects. In the meta-analysis, as the eligible studies used a variety of different populations, a random effects model (rather than a fixed effects model) was determined to be more suitable. In the case of 1 included literature, the p-value of heterogeneity was 1. In addition, the effect size of the fixed effect and the random effect model is the same. The statistical difference between subgroups was evaluated with a meta-regression test. We used Begg’s funnel plot and Egger’s test to assess the publication bias; if significant publication bias was found, the fail-safe N and trim-fill tests were additionally used to confirm the degree of publication bias. The results were considered statistically significant at p < 0.05.The meta-analysis was performed using the Comprehensive Meta-Analysis software package . The rates of EBV-infected GCs were evaluated in eligible studies, and the differences in overall survival between EBV-infected and non-infected GCs were investigated. Subgroup analyses for EBV positivity rates were performed based on study and tumor locations. In survival analysis, subgroup analyses were performed based on Lauren’s classification and molecular classification. In addition, the overall survival between patients with EBV-infected GCLS and those with non-GCLS were compared. We checked the heterogeneity between the studies based on the Q and In = 256), articles on other diseases (n = 195), written in a language other than English (n = 46), and duplication (n = 2). Finally, 57 eligible articles were selected and included in the meta-analysis . In the subgroup analysis based on study location, the EBV positivity rates in the Asian, American, and European subgroups were 0.114 (95% CI 0.080–0.158), 0.107 (95% CI 0.079–0.142), and 0.084 (95% CI 0.070–0.100), respectively. In the subgroup analysis, there was no significant publication bias. The EBV positive rates of antrum, body, and cardia/fundus were 0.083 (95% CI 0.055–0.123), 0.130 (95% CI 0.095–0.176), and 0.182 (95% CI 0.126–0.257), respectively.First, the EBV positive rates of GCs were investigated and analyzed through a meta-analysis. The estimated EBV positivity rate of GCs was 0.104 . Next, a subgroup analysis was performed based on the histologic subtype, Lauren’s classification, and molecular classification. There was a significant difference in the overall survival between the patients with EBVaGC and those with non-EBVaGC in GCLS, but not in non-GCLS . The Egger’s test was not performed due to the number of included articles in GCLS and non-GCLS subgroups. In the diffuse type of Lauren’s classification, the patients with EBVaGC had better prognosis than those with non-EBVaGC . However, there was no significant difference in the overall survival between the patients with EBVaGC and those with non-EBVaGC in the intestinal type of Lauren’s classification . In addition, Egger’s test was not performed due to the number of included articles in intestinal and diffuse type subgroups. In the molecular classification, there were no significant differences in the overall survival between the patients with GCs who were EBV positive and MSI-high (MSI-H) and those with GCs who were microsatellite stable (MSS) and EBV negative . There was no significant publication bias in this subgroup analysis. In the Asian and American subgroups, the patients with EBVaGC showed better prognosis than those with non-EBVaGC . We additionally performed a subgroup analysis based on various factors such as study location, Lauren’s classification, and molecular classification. In the subgroup analysis based on the study location, statistical significance was found in the Asian and American subgroups, but not in the European subgroup. This result was similar to that of a previous meta-analysis . AccordiThe different tumor behaviors of GCs can be affected by their histologic and molecular characteristics ,66, and The patients with EBVaGC have better prognosis compared with those with non-EBVaGC. Similar results were obtained in our meta-analysis. However, individual studies reported different results, while some showed contrasting results. The intestinal type of EBVaGC had a worse prognosis than the intestinal type of non-EBVaGC . In addiThis study has some limitations. First, some studies show that EBVaGC is associated with worse prognosis. This discrepancy might be due to the different populations evaluated and methods used for detecting EBV. The significance of the prognostic implication of EBV was reported in 7 out of 32 Asian studies, 1 out of 11 American studies, and 1 out of 12 European studies. The individual studies had sample sizes ranging from 43 to 1328 participants. The EBV positivity ranged from 2.0% to 40.1%. As the individual studies included a variety of different populations, the results may vary between eligible studies. However, our meta-analysis did not confirm this cause. Second, differences were found in the number of studies conducted before and after the introduction of the molecular classification. The amount of analyzed results according to the molecular classification were too few compared with the total number of analyzed results. Third, the patients with GCLS and those with diffuse-type GC and EBV infection had better prognosis. However, since the number of studies presenting this information is too small, further studies are warranted. Fourth, the EBV positive rates were differed by the tumor location. However, we could not perform the detailed analysis for the prognostic impacts of EBV based on the tumor location due to insufficient information. Further study on the impact of the tumor location will be useful for predicting the prognostic role of EBV.Although EBV infection was significantly correlated with favorable survival in the patients with GCs, the details of this patient group differed according to the new molecular classification and by subtype. However, this meta-analysis may still be considered significant as the prognostic role of EBV may vary depending on the GC subtype. Hence, further detailed studies on the subtypes with fewer cases are needed.


Oral lichenoid conditions are common potentially malignant disorders affecting oral mucosa. The 2003 modified WHO criteria is the most widely used criteria in diagnosing oral lichen planus; however, the concern on the implementation of epithelial dysplasia as an exclusion criterion may result in the remarkable underestimation of the potential malignancy of this disease continuing. This systemic review and meta-analysis attempted to provide an objective view by comparing the malignant transformation rate of oral lichen planus diagnosed according to the 2003 modified WHO criteria and non-2003 criteria, and to determine risk variables associated with malignant transformation of oral lichenoid conditions. We found that the malignant transformation rate of oral lichen planus differed based on diagnostic criteria, with lower rates for the 2003 modified WHO criteria than non-2003 criteria, albeit not statistically significant. A higher incidence of malignant transformation was found for red-type lesions and patients who are smokers, alcohol consumers, and HCV positive.Objectives: This systematic review and meta-analysis aims to evaluate the evidence on the malignant potential of oral lichenoid conditions (OLCs) including oral lichen planus (OLP), oral lichenoid lesions (OLL), and lichenoid mucositis dysplasia (LMD). In addition, it aims to compare the rate of malignant transformation (MT) in OLP patients diagnosed according to different diagnostic criteria, and to investigate the possible risk factors for OLP MT into OSCC. Materials and methods: A standardized search strategy was applied across four databases . Screening, identification and reporting followed the PRISMA framework. Data on MT were calculated as a pooled proportion (PP), subgroup analyses and possible risk factors for MT were pooled as odds ratios (ORs). Results: Among 54 studies with 24,277 patients, the PP for OLCs MT was 1.07% . The estimated MT rate for OLP, OLL and LMD was 0.94%, 1.95% and 6.31%, respectively. The PP OLP MT rate using the 2003 modified WHO criteria group was lower than that using the non-2003 criteria . A higher odds ratio of MT was observed for red OLP lesions , smokers , alcohol consumers  and those infected with HCV , compared to those without these risk factors. Conclusions: OLP and OLL carry a low risk of developing OSCC. MT rates differed based on diagnostic criteria. A higher odds ratio of MT was observed among red OLP lesions, smokers, alcohol consumers, and HCV-positive patients. These findings have implications for practice and policies. Oral lichen planus (OLP) is a T-cell-mediated chronic inflammatory disorder affecting the oral mucosa . Based oMoreover, it is still controversial if LMD is a distinct histopathologic entity. Farah et al. recommended that LMD should not be classed as a distinct pathological entity based on their finding that the transcriptomic and immunophenotypic profiles of LMD were similar to that of OLP but distinct from oral epithelial dysplasia . In 2020A number of studies have investigated the malignant potential of OLCs since the first case report of oral squamous cell carcinoma (OSCC) developed from OLP 1910 . The repThe objective of this systematic review and meta-analysis was to investigate the malignant potential of OLCs, to compare the rate of MT of OLP based on the diagnostic criteria of the 2003 modified WHO criteria and non-2003 modified WHO criteria, and to determine risk variables associated with MT of OLCs.The systematic review and meta-analysis were performed according to the Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) and Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelines.PubMed, Embase, Web of Science and Scopus were searched to identify eligible studies. Title, abstract and keywords of studies were searched using free terms and combined database thesaurus terms including Medical Subject Headings (MeSH) and Embase Subject Headings (Emtree) to maximize sensitivity . Two revInclusion criteria: (1) Full-length, peer-reviewed original research papers published in English from January 2003 to September 2022; (2) Observational studies providing sufficient information on the diagnosis and rate of MT of OLCs.Exclusion criteria: (1) Letters to the editor, case reports, reviews, book chapters and any study in a language other than English; (2) Studies that do not differentiate between OLP and cutaneous LP or lesions from other anatomical sites; (3) In vitro or animal studies, and cross-sectional clinical studies without providing the data of follow-up and MT.The following data from each included study were extracted: (1) year and region of the study, study design, type of lesions and follow-up details including range, mean, median and frequency; (2) risk of bias assessment domains; (3) diagnostic criteria; (4) clinical patterns of OLP including ; (5) characteristics of MT occurrence including: c. location; b. developed from which type of OLCs  or sub-type of OLP (white or red lesions); (6) associated risk factors of MT; (7) detailed descriptions of confirmed OLP/OLL/LMD-related malignancies. Disagreements between reviewers were resolved by consensus or by the decision of a third independent reviewer.Newcastle–Ottawa scale was used to assess the risk of bias within the included observational studies with the following modifications:Selection: (1) A cohort of patients that is representative of the general population; (2) the diagnosis of OLCs must be based on both clinical and histopathological confirmation; (3) cases of OLP/OLL and epithelial dysplasia must be well differentiated.Comparability of groups on demographic characteristics and important potential confounders: While presenting variables from the study cohort, risk factors including smoking and alcohol consumption must be documented.Outcomes: (1) Independent or blind assessment stated in the paper; (2) all variables associated with MT must be documented in details including age, gender, clinical sub-type, location of lesions and tumors, follow-up information, TNM, related risk factors and intervention; (3) sufficient follow-up duration (2 years or more); (4) adequacy of follow-up of cohorts.In view of the design of the studies included without non-exposed group, we modified the Newcastle–Ottawa Scale to suit the selected studies by eliminating question No.2 (the selection of the non-exposed group). In this case, the total Newcastle–Ottawa Scale has been revised to be out of eight, rather than nine . Each stThe quality of evidence was evaluated using the Grading of Recommendations Assessment, Development and Evaluation (GRADE) system. The quality of evidence was classified in one of four levels: very low, low, moderate or high. The overall quality rating was based on the following domains: risk of bias, inconsistency, indirectness, imprecision and publication bias .p < 0.05 was considered significant. Cumulative meta-analysis was performed calculating the pooled proportion of the overall risk of MT in OLCs and the subtype of OLP, OLL and LMD under the random effects model using Der-Simonian Liard method. Heterogeneity was evaluated using Cochran’s Q statistic and quantified using Higgin’s I2 statistics. Subgroup analysis was conducted to compare the difference in MT based on the diagnostic criteria, regional disparity and location of OLP. To investigate the possible risk factors that could potentially lead to the malignization of OLCs, we performed a comprehensive analysis based on gender of patients, clinical sub-type, habits of smoking or alcohol consumption, infection with hepatitis C virus (HCV), diabetes mellitus and hypertension. The odds ratio (OR), along with 95% confidence interval (CI), was applied for the representation of pooled results from studies. Publication bias was assessed using a funnel plot and quantified using Egger’s linear regression test. Asymmetry of the collected studies’ distribution by visual inspection or p-value < 0.1 indicated publication bias ) a,b.p = 0.241; p < 0.001; p = 0.043; p = 0.031; 2 = 0%) with 6850 patients revealed a significantly higher risk of MT in HCV-positive patients in comparison to the HCV-negative patients ; however, the difference was not significant . Full details of the risk of bias assessment are shown in This systematic review revealed an overall MT rate of 1.07% in OLCs based on the analysis of 54 studies with a total of 24,277 patients. Overall, 0.94% of OLP patients developed OSCC, which represents the lowest risk among the OLCs, while the MT rate of OLL and LMD was 1.95% and 6.31%, respectively.p = 0.667). To our knowledge, this is the first study comparing the MT rate of OLP patients diagnosed with different diagnostic criteria. It is worth noting that our meta-analysis revealed that the MT rate of LMD was significantly higher than that of OLP (p = 0.023) Our findings advocate the suggestion from González-Moles et al.: “Further studies and a revision of the diagnostic criteria of the disease are required to establish the true prevalence of this important risk marker for cancer progression” [The presence of epithelial dysplasia is considered as the gold-standard indicator for the assessment of cancer risk and progression in OPMD. Whether epithelial dysplasia demonstrates OLP/OLL-like features, or OLP/OLL lesions gain dysplastic features during the progress of the disease, is a continuing debate. Some clinical and pathological evidence, including our recent study, revealed that OLP may develop epithelial dysplasia along the course of its evolution. In a recent case study, we found that one patient with OLP  developed well-differentiated OSCC and moderate epithelial dysplasia at two different sites of long-standing lichen planus at the tongue. This finding supports the concept that OLP may develop dysplastic features during the progress of the disease, which may finally progress to OSCC. While solid evidence is still lacking, many clinicians and pathologists raised the issue of whether or not the detection of epithelial dysplasia invalidates the diagnosis of OLP. The poorly understood pathogenesis of OLP leads to the inconsistency in diagnostic criteria. While the 2003 modified WHO criteria have become the current consensus for the diagnosis of OLP, the consideration of epithelial dysplasia as an exclusion criterion may result in a remarkable underestimation of the potential malignancy of this disease . Gonzáleression” .p = 0.728).While OLP and OLL tend to be considered as distinct conditions in the diagnostic criteria proposed by van der Meij and van der Waal, they do share significant overlapping features ,80. In ap < 0.001). The upregulation of the expression of p53 and matrix metalloproteinases, the potential cancer biomarkers in red-type OLP, provides basic evidence of mechanism of the cancer development [To investigate the risk factors that could potentially prompt the malignization of OLP, we conducted a comprehensive analysis on the variables including the gender, clinical patterns, smoking and alcohol consumption, HCV, diabetics and hypertension based on the available data. Aghbari et al. reported that erosive-type OLP carries the highest risk of MT (1.7%), followed by atrophic (1.3%) and reticular (0.1%) patterns . Lanioszelopment ,83.smoke = 1.79, Psmoke = 0.043) and alcohol users  compared to those patients without the habits. Smoking is categorized as a major risk factor of oral cancer. It is highly debatable whether smoking independently causes OSCC or just interacts with OLP to increase its malignant potential in the OLP-OSCC patient. Therefore, as part of the 2003 WHO modified criteria, smoking documentation must be in placed in order to identify the true MT. Alcohol may promote malignancy through atrophying the oral epithelium and increasing the permeability of epithelial cells. Chronic use of alcohol has been reported to have genotoxic and mutagenic effects and to impair both innate and acquired immunity, increasing the risk of infections and cancers [Consistently with some other studies ,84, our  cancers . Meanwhi cancers ,87.p < 0.001). Rangel et al. reported that 12.8% of head and neck cancer patients were HCV positive [Another risk factor for OLP malignization to emerge in this meta-analysis was the presence of HCV infection. HCV-positive patients showed significantly higher MT prevalence compared to the HCV-negative patients (positive . Severalpositive ,90. Nagapositive . Moreovepositive . Howeverpositive ,93. ThisSystemic diseases such as DM and hypertension are known to increase the risk of significant oral problems, which are often detected in cases of OLP ,75,94. TTo the best of our knowledge, this is the most comprehensive systematic review and meta-analysis to focus on the MT rate and potential risk factors of OLCs. We attempt to overcome the potential diagnostic confusion on OLP, OLL and LMD by categorizing the cases into different analytic datasets to calculate the MT of each pathological condition independently.It has been 20 years since the 2003 modified WHO criteria was Introduced, and they have become the most widely used criteria in diagnosing OLP. While concerns remain that the implementation of epithelial dysplasia as an exclusion criterion may result in a remarkable underestimation of the potential malignancy of this disease, we attempt to provide an objective view by comparing the MT rate of OLP patients diagnosed according to the 2003 modified WHO criteria and non-2003 criteria. The result demonstrates that the MT rate calculated from the OLP patients diagnosed with the criteria proposed by van der Meij and van der Waal is lower than that with the non-2003 criteria, but the difference was not significant.There are several potential limitations associated with our study. Firstly, we only included studies published in the English language. Secondly, despite our efforts, it may not be fully possible to distinguish OLP, OLL and LMD in the included studies. Third, not all studies provided numerical data of associated risk factors, which may limit the implementation of our meta-analysis in identifying the role of MT in OLP. Finally, the follow-up periods among the included studies were highly variable, leaving the possibility that some OLP/OLL patients would have progressed to OSCC if followed up for a longer period.Ideally, to investigate the MT risk of oral potentially malignant disorders is to conduct a prospective clinical trial with a regular and long follow-up. However, practically speaking, this is very difficult due to the highly variable risk and lengthy time interval of the transition of premalignancy to OSCC. Considering the difficulties in conducting a prospective clinical trial, a large-scale multicenter cohort study would be the optimal option to address this issue. Proper clinical and histopathological documentation, including careful differentiation between OLP and other lichenoid lesions, and detailed numerical data of associated risk factors, should be emphasized in future research. As the OLP is considered to be premalignant, an evidence-based recall system would be useful to improve the chance of early diagnosis of oral cancer; however, the evidence regarding the frequency of follow-up/recall is very small and far from reliable due to controversy on the risk of MT. The majority of the reported OPMD risk assessment models are based on conventional statistical approaches. Machine learning algorithms represent an advancement to conventional statistical models for evaluating the risk of MT. The deep learning algorithm may not only provide an effective and reliable prediction and decision-making model for clinicians to accurately estimate the risk of MT, but it also builds a solid ground for future development and implementation of a user-friendly web-based platform not only for specialists but also for dentists and family doctors planning management and monitoring protocols. In addition, efficacious communication with patients is also vitally important to raise public awareness about the malignant potential of the oral lichenoid condition.The systemic review and meta-analysis provide evidence that OLP and OLL carries a low risk of developing OSCC. The MT rate of OLP differed based on diagnostic criteria, with lower MT rates for the 2003 modified WHO criteria than non-2003 criteria, albeit not statistically significant. A higher incidence of MT was found for red OLP lesions and patients who are smokers, alcohol consumers, and HCV positive. The mechanism dictating how these risk factors promote MT needs further clarification. An evidence-based recall system would be useful to improve the chance of early diagnosis of oral cancer, but such a system is not yet available. The deep learning algorithm and web-based platform may provide a reliable and user-friendly prediction and decision-making model not for only specialists but also for dentists and family doctors planning management and monitoring protocols.


Calcifications on mammography can be indicative of breast cancer, but the prognostic value of their appearance remains unclear. This systematic review and meta-analysis aimed to evaluate the association between mammographic calcification morphology descriptors (CMDs) and clinicopathological factors.A comprehensive literature search in Medline via Ovid, Embase.com, and Web of Science was conducted for articles published between 2000 and January 2022 that assessed the relationship between CMDs and clinicopathological factors, excluding case reports and review articles. The risk of bias and overall quality of evidence were evaluated using the QUIPS tool and GRADE. A random-effects model was used to synthesize the extracted data. This systematic review is reported according to the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA).Among the 4715 articles reviewed, 29 met the inclusion criteria, reporting on 17 different clinicopathological factors in relation to CMDs. Heterogeneity between studies was present and the overall risk of bias was high, primarily due to small, inadequately described study populations. Meta-analysis demonstrated significant associations between fine linear calcifications and high-grade DCIS , (comedo)necrosis , (micro)invasion , and a negative association with estrogen receptor positivity .CMDs detected on mammography have prognostic value, but there is a high level of bias and variability between current studies. In order for CMDs to achieve clinical utility, standardization in reporting of CMDs is necessary.Mammographic calcification morphology descriptors (CMDs) have prognostic value, but in order for CMDs to achieve clinical utility, standardization in reporting of CMDs is necessary.CRD42022341599• Mammographic calcifications can be indicative of breast cancer.• The prognostic value of mammographic calcifications is still unclear.• Specific mammographic calcification morphologies are related to lesion aggressiveness.• Variability between studies necessitates standardization in calcification evaluation to achieve clinical utility.The online version contains supplementary material available at 10.1186/s13244-023-01529-z. Screening mammograms frequently reveal the presence of calcifications, a majority of which are associated with benign changes in breast tissue. However, a minority of these calcifications may be linked to invasive breast cancer (IBC) or its non-obligatory precursor, ductal carcinoma in situ (DCIS) .The introduction of systematic mammographic screening –3 has siGiven the heterogeneity of DCIS in terms of morphology, biology, genetics, and outcome , mammogrThe American College of Radiology’s Breast Imaging-Reporting and Data System (BI-RADS) standardizes these patterns, classifying calcifications with a suspicious morphology into four categories, (1) amorphous, (2) coarse heterogeneous, (3) fine pleomorphic, and (4) fine linear or fine linear branching . It is nInvestigating the association between calcification morphology descriptors (CMDs) and specific clinicopathological factors of DCIS lesions, such as grade, receptor-based surrogate subtypes based on hormone receptors and HER2, and the risk of local DCIS or IBC recurrence could provide clinicians with valuable insights into the likelihood of DCIS progression and lesion aggressiveness. Accurately determined CMDs may enable clinicians to make better-informed decisions regarding the necessity of further diagnostic procedures, such as biopsies. However, despite numerous studies, the prognostic value of mammographic calcification descriptors remains unclear –48.To address this knowledge gap, we conducted a systematic review and meta-analysis to assess the association between mammographic CMDs and clinicopathological factors in women with DCIS, while evaluating the overall quality of evidence and identifying sources of bias.This systematic review is reported according to the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) . The stuA comprehensive literature search was performed in Medline via Ovid, Embase.com, and Web of Science Core collection  according to Bramer et al. . Non-peeThe search results were imported into EndNote 20  to removNo other sources were searched, and no other methods were used.After the selection process, a custom form was used to extract several study characteristics of the included articles. These characteristics included study details , patient and lesion characteristics , age in years, histopathological size of the lesion), outcome measurement details (type of assay), exposure measurement details  readers, calcification classification system), and necessary information for quality assessment.The numerical results were documented by cross-tabulating each CMD’s absolute numbers concerning clinicopathological factors.Authors were not contacted for missing data. Missing information was noted as “not specified” or “not available”.The Quality in Prognostic Studies (QUIPS) tool  was usedTo ensure consistency in the interpretation of the QUIPS criteria, the first five papers’ results were compared between the two reviewers. Then, M.M.L. and S.D. independently performed quality assessment of the included studies. A low-high discrepancy was defined as a difference in risk of bias rating between the two reviewers for a specific domain. When the reviewers did not reach a consensus in case of a low-high discrepancy, they sought the help of a third reviewer (A.W.B.D.), to reach a consensus decision.t-test for studies published between 2000 and 2010 and between 2010 and 2022, with high risk of bias scored as three, medium risk as two, and low risk as one.The average risk of bias per QUIPS domain was compared using a The certainty of evidence (CoE) for studies included in the meta-analysis was assessed using a modified GRADE  instrument . Since aI2 values and the Cochran’s Q-test’s p-value (P(Q)). I2 values above 50% and a significant Q-test indicated high heterogeneity between the studies. Indirectness was assessed by examining differences between studies in population characteristics, exposure, and outcome measurements. Imprecision was evaluated based on point estimates spread, 95% confidence intervals size, and the overlap of confidence intervals. Publication bias was assessed using the Egger’s test. To account for the test’s lack of power without a representative number of studies [The risk of bias was determined using the average QUIPS tool score. Inconsistency (heterogeneity) was determined through  studies , it was The total number of studies assessing the same clinicopathological factor in relation to CMDs was cross-tabulated along with the number of studies finding a significant association with CMDs.The associations between CMDs and clinicopathological factors were synthesized in pooled odds ratios (pORs) after grouping CMDs into three risk categories. CMDs were categorized into a low-risk, intermediate-risk, and high-risk group for meta-analysis since not all studies used BI-RADS descriptors or analyzed different CMDs separately. The low-risk group included calcifications with punctate or amorphous morphology and served as the reference group. The intermediate-risk group consisted of calcifications with a coarse heterogeneous or (fine) pleomorphic morphology, while the high risk-group included fine linear calcifications based on the difference in positive predictive value for the presence of DCIS. To categorize CMDs into risk groups, non-BI-RADS descriptors were aligned with BI-RADS descriptors (Supplementary Table Stau2) in calculating the pORs, as this estimator is suitable for studies with small sample sizes and binary outcomes. A forest plot was used to visually represent the pOR for each factor and risk group.Random-effects models were employed to calculate the pORs for CMDs and clinicopathological factors, allowing for heterogeneity between studies. The Mantel-Haenszel method was used to combine binary effect estimates (ORs) across studies. The Paule-Mandel estimator was used to model between-study variance  that were not comparable with the other studies assessing that specific clinicopathological factor.Subgroup analyses were not feasible due to the limited number of studies per clinicopathological factor and insufficient information on relevant subgroups .p-values of < 0.05 were used to determine statistical significance.All analyses were performed with R version 4.2.2  using the meta R package (version 6.1). Pooled estimates were reported in combination with 95% confidence intervals (CI), and two-sided Through an extensive search that included reference cross-checking of relevant articles, a total of 4715 unique articles were collected. An initial screening, based only on studies’ titles, led to the exclusion of 3946 articles. A following evaluation of the remaining 769 articles, based on their abstract, led to the removal of 666 articles. Among these, about 44% were excluded because they were not in English, presented non-original data, or were classified as case reports. Further analysis of the remaining 103 articles using full-text assessment  study characteristics, presenting an overview of the included studies and patient populations; (ii) results of synthesis, consolidating the extracted outcomes from the included studies; and (iii) assessment of bias using the QUIPS tool, evaluating the risk of bias across the selected studies.Figure n = 27), case–control (n = 1), and a case-cohort within a random control trial (n = 1) designs were used. Twenty were single-center studies, and 9 involved multiple centers. While 20 studies exclusively studied DCIS, nine studies included patients with both DCIS and IBC. The number of DCIS patients with calcifications per study ranged from 32 to 1783. Fifteen studies described calcification morphology according to the BI-RADS system, while the remaining 14 studies used non-BI-RADS descriptors. In ten studies, the lesions were described as only screen-detected, while in another ten studies, they were reported to be both screen-detected and non-screen-detected. The remaining studies did not specify the method of lesion detection. Thirteen studies specified using mammograms with calcifications only, without other mammographic abnormalities.In all 29 studies, the data was collected retrospectively from hospital registries, national registries, or clinical trials. Cohort (Table n = 16), (micro)invasion (n = 8), (comedo)necrosis (n = 7), HER2 overexpression (n = 6), ER positivity (n = 6), age (n = 3), Ki67 or proliferation (n = 2), histological size (n = 2), neoductgenesis (n = 2), calcification distribution (n = 2), margin status (n = 1), comedocarcinoma (n = 1), multicentricity (n = 1), tenascin-C (n = 1), and Oncotype DX score (n = 1). Furthermore, five studies assessed prognostic outcomes including recurrence (n = 4) and DCIS progression to IBC (n = 1).A total of 29 studies investigated 17 distinct factors concerning CMDs (Table n = 11), HER2 overexpression (n = 4), ER positivity (n = 4), (comedo)necrosis (n = 5), and the presence of (micro)invasion) (n = 5). The meta-analysis shows the aggregated results for low-, intermediate-, and high-risk CMDs concerning the clinicopathological factors.Out of the 17 clinicopathological factors reported across 29 studies, 14 factors were significantly association in at least one study , (comedo)necrosis , (micro)invasion , and ER positivity . High-risk CMDs were negatively associated with ER positivity, indicating a reduced incidence of ER positivity in high-risk versus low-risk CMDs.p-value of 0.09 for (micro)invasion.Intermediate-risk CMDs were significantly associated with high grade  and (comedo)necrosis , while showing an increased pOR of 1.66  with a I2 and P(Q) revealed inconsistencies in the estimates reported in the included studies concerning high grade : 54%, p = 0.002 for high-risk and 47%, p = 0.04 for intermediate-risk CMDs), ER positivity : 49%, p = 0.12 for high-risk CMDs), comedo(necrosis) : 52%, p = 0.08 for high-risk CMDs), and invasion : 55%, p = 0.06 for intermediate-risk CMDs).Heterogeneity measures I2 with > 83% and p(Q) <  = 0.001.Neither high-risk CMDs  nor intermediate-risk CMDs  were significantly associated with HER2. One contributing study by Zhou et al.  reportedEgger’s test was not significant, indicating that there was no publication bias for high grade, while publication bias was not determined due to the small sample sizes for the other outcomes.According to the GRADE tool approach, the certainty of evidence for all outcomes can be rated as low , these domains were assigned a moderate risk of bias rating. This suggests that the discrepancies in the assessment of the bias using the QUIPS tool were limited.The study participation domain revealed eight studies with a high risk of bias and 15 with a moderate risk in either prognostic or non-prognostic outcomes. The downgrading of studies was primarily attributed to small sample sizes and inadequate description of study groups, data collection criteria, and methods or reasons for missing data.In the exposure domain, seven studies exhibited a high risk of bias, while 13 demonstrated a moderate risk. The downgrading mainly resulted from situations where only one reader determined the CMDs, or when crucial details were omitted, such as whether the readers were blinded to the outcome and how consensus was achieved between readers.The outcome measurement domain indicated three studies with a high risk of bias and 11 studies with a moderate risk. High-risk studies were characterized by a severe lack of detail regarding the definition and method of measuring the outcome variable. Moderate-risk studies contained insufficient information on the measurement of outcome variables and, if applicable, blinding of reviewers.With regard to the confounding domain, most individual studies did not adjust their results for potential confounders. Five studies were rated as having a high risk of bias in this domain and 18 as having a moderate risk of bias. High-risk studies failed to account for potential confounding through matching, stratification, or the initial assembly of comparable groups. Prognostic studies were rated as moderate or high risk when they did not adjust for treatment or age in their statistical analyses. Studies with a design that somewhat limited the risk of confounding were rated as moderate risk.The statistical analysis and reporting domain predominantly displayed a low risk of bias. However, in thirteen studies, this domain was rated as moderate, because the analysis was not powerful enough to prove or disprove the hypothesis. This occasionally occurred for individual CMD groups, e.g., when chi-square tests were applied to small sample sizes.In the study attrition domain, three prognostic studies were rated as having a moderate risk of bias because the follow-up or characteristics of women who completed the study and those who did not were not described.p = 0.01) and confounding (p = 0.025) domains for studies published before 2010 compared to those published after 2010.Notably, the average risk of bias was significantly higher in the exposure measurement  and clinicopathological factors in women diagnosed with DCIS with the aim of distinguishing high-risk from low-risk DCIS lesions based on CMDs, which may aid clinical decision making. A total of twenty-nine studies were identified, evaluating 17 clinicopathological factors, of which five were deemed appropriate for meta-analysis.The meta-analysis revealed a significant association between fine linear calcifications, i.e., the high-risk group and features of aggressiveness including high grade, presence of (comedo)necrosis, and (micro)invasion. An inverse association was observed with ER positivity.Intermediate-risk CMDs, i.e., coarse heterogeneous and fine pleomorphic calcifications, were significantly associated with high-grade and (comedo)necrosis. The associations were generally similar to those of the fine linear calcifications, albeit to a lesser extent. The presence of calcifications in DCIS is thought to be due to active secretion of calcium into the ducts by  epithelial cells in non-comedocarcinoma and calcification of necrotic debris in comedocarcinoma . The obsWith respect to HER2 overexpression, a positive association with high-risk CMDs was found, whereas a negative association was identified with intermediate-risk CMDs; however, neither association was statistically significant. The results were inconsistent, due to one study  presentiConsidering the prior findings that ER-positive and HER2-negative  breast cancers are generally less aggressive than ER-negative and HER2-positive invasive breast cancers , the assTwelve other clinicopathological factors were examined in relation to CMD, in addition to the ones that were suitable for inclusion in the meta-analysis, and several studies reported either significant or non-significant associations for these factors. Among these factors, the one related to the risk of recurrence including the Oncotype DX score showed significant association in two out of five studies. However, additional cohorts and standardized methods are required to validate the evidence on these factors.Our comprehensive analysis suggests a potential association between CMDs and the aggressiveness of lesions, particularly in the progression from DCIS to IBC. This conclusion is supported by O’Grady et al. , and TotUncertainty in the findings of the meta-analysis were revealed and the most common sources of bias in the relevant articles using the GRADE approach and the QUIPS tool. The primary source of bias originated from the study participation domain due to low sample size and inadequately described study groups. The majority of the included studies utilized data from hospital or national registries, which could have influenced the found estimates. Retrospective registry-based studies depend on the quality, size, and completeness of relevant variables and features of the used registries. Frequently, these variables were not mentioned or properly described.The exposure domain was also frequently rated as moderate or high risk of bias because CMD determination was often conducted by a single reader, given that the assessment of these qualitative descriptors is prone to inter- and intra-observer variability. While radiologists strongly agree on the presence of calcifications, they agree to a lesser degree on the classification of the observed calcification morphology . FurtherBias due to inter-reader variability could also have occurred in the outcome domain for the clinicopathological factor grade, meaning that both radiomics and pathomics methods might improve risk stratification of calcifications.Concerning the confounding domain, most studies reported on the morphology of calcifications in isolation, not considering other descriptors that can aid in further risk stratification and control for confounding factors such as distribution, size, and clinicopathological factors. Some studies also assessed distribution, as this is another calcification descriptor often used in clinical practice, but not in combination with morphology. Hence, the results from this review reflect univariable associations only, which can lead to biased estimates and incorrect conclusions if relevant covariates are omitted.The studies published after 2010 had significantly lower risk of bias scores in the exposure and confounding domains as compared to the older studies from before 2010, indicating an improvement in study and evaluation methods for this research question.As with the majority of systematic reviews, the design of the current study is subject to potential limitations . SystemaStudies with a high risk of bias according to the QUIPS tool were not excluded in our meta-analysis given that the risk of bias was relatively high for all studies, predominantly due to small sample sizes and inconsistencies in the evaluation and registration of calcifications. These biases have affected the reported pOR estimates, as well as the performance of tests of heterogeneity, publication bias, and other sample size effects summarized in the GRADE score for each clinicopathological factor and calcification group. For a more nuanced interpretation of the meta-analysis results, the ORs and QUIPS scores per domain for each study were reported.Nonetheless, this is the first systematic review and meta-analysis to assess the association between mammographic CMDs and clinicopathologic factors in women with DCIS. The primary strength of such a meta-analysis lies in its capacity to enhance the identification of associations and uncover the sources of heterogeneity between reported estimates across studies. Indeed, using the QUIPS tool, we identified the most frequently occurring biases in included studies by assessing the association between CMDs and clinicopathological factors in a standardized manner.In conclusion, our meta-analysis demonstrates that specific mammographic calcification morphologies are related to clinicopathological factors associated with lesion aggressiveness in women with DCIS.This systematic review also showed a high risk of bias and heterogeneity between studies. Therefore, these findings need to be verified through high-quality studies that use homogeneous cohorts and standardized, reliable calcification assessment systems. Future radiomics and deep learning studies may help in the extraction of relevant calcification features that can extract prognostic information in DCIS lesions and, ultimately, in making the distinction between high-risk and low-risk DCIS lesions and reducing overtreatment of DCIS.Additional file 1: Supplementary Table S1. S1a Table: Search strategy used in Medline via Ovid. S1b Table: Search strategy used in Embase. S1c Table: Search strategy used in Web of Science.Additional file 2: Supplementary Table S2. QUIPS master form.Additional file 3: Supplementary Table S3. Grade assessment.Additional file 4: Supplementary Table S4. Alignment of BIRADS and Non-BIRADS calcification morphology descriptors.Additional file 5: Supplementary Table S5. Excluded articles after full text screening.


To estimate the dose–response associations between non-occupational physical activity and several chronic disease and mortality outcomes in the general adult population.Systematic review and cohort-level dose-response meta-analysis.PubMed, Scopus, Web of Science and reference lists of published studies.Prospective cohort studies with (1) general population samples >10 000 adults, (2) ≥3 physical activity categories, and (3) risk measures and CIs for all-cause mortality or incident total cardiovascular disease, coronary heart disease, stroke, heart failure, total cancer and site-specific cancers .196 articles were included, covering 94 cohorts with >30 million participants. The evidence base was largest for all-cause mortality , and incidence of cardiovascular disease  and cancer . In general, higher activity levels were associated with lower risk of all outcomes. Differences in risk were greater between 0 and 8.75 marginal metabolic equivalent of task-hours per week (mMET-hours/week) , with smaller marginal differences in risk above this level to 17.5 mMET-hours/week, beyond which additional differences were small and uncertain. Associations were stronger for all-cause (relative risk (RR) at 8.75 mMET-hours/week: 0.69, 95% CI 0.65 to 0.73) and cardiovascular disease  mortality than for cancer mortality . If all insufficiently active individuals had achieved 8.75 mMET-hours/week, 15.7% (95% CI 13.1 to 18.2) of all premature deaths would have been averted.Inverse non-linear dose–response associations suggest substantial protection against a range of chronic disease outcomes from small increases in non-occupational physical activity in inactive adults.PROSPERO registration number CRD42018095481. Cardiovascular disease (CVD) is the leading cause of death globally, responsible for 17.9 million annual deaths in 2019,Higher levels of physical activity (PA) are associated with lower risk of all-cause mortality, CVD and some site-specific cancers.Accurate estimation of the dose–response association between PA and disease outcomes, combined with prevalence estimates, is a prerequisite to assessing the population disease burden of insufficient PA and the potential impact of changes to population levels of PA. Modelling studies that evaluate public health strategies aimed at increasing population activity levels have found that the shape of the dose–response association makes a significant difference when quantifying population health impacts.Our objective was to quantify dose–response associations between non-occupational PA and several CVD, cancer and mortality outcomes using a novel harmonisation framework to overcome the challenges posed by different PA measurement methods. This framework allowed us to compare studies measuring and reporting PA in a wide variety of ways on the same activity exposure scale of marginal MET-hours per week (mMET-hours/week). This meant we could be more inclusive of the available evidence than previous meta-analyses and we could explore the dose–response relationships between non-occupational PA and nine site-specific cancers for the first time. To contextualise our results for public health promotion, we generated potential population impact fractions (PIFs) to estimate the proportion of preventable deaths and disease outcomes at different non-occupational PA levels.The study protocol is available at PROSPERO (registration number CRD42018095481).et al.10.1136/bjsports-2022-105669.supp1Supplementary dataOnly articles that examined leisure-time PA, alone or in combination with other domains or specific types of activity, were included. We excluded articles whose measures of PA contained occupational activity that could not be factored out and that investigated individual domains of PA that did not include leisure-time activity .The outcomes of interest were the following:All-cause, total CVD and total cancer mortality.Total CVD, coronary heart disease, stroke and heart failure incidence .Total and site-specific  cancer incidence . These site-specific cancers were selected based on previously reported associations with PA.We searched PubMed, Scopus, Web of Science and the reference lists of reviews retrieved from our systematic review, known to the authors or cited in the 2018 US Physical Activity Guidelines Advisory Committee Scientific Report.Titles and abstracts, and subsequently full texts, were screened independently twice for eligibility . DisagreIf multiple articles reported results on the same cohort and outcome, we followed specific criteria  to selecData from each paper were extracted by one reviewer and double-checked independently by a second reviewer. Disagreements were checked against the full text and resolved by discussion.We extracted data on publication , study characteristics , PA exposure assessment  and outcome assessment  codes as reported in the PROSPERO registration and For each exposure category, we extracted information to quantify PA volume, number of cases, number of participants and/or person-years of follow-up, and risk estimates with 95% CIs. Risk estimates from the most adjusted model were used. When available, risk estimates of the most adjusted model without adiposity-related covariates , waist circumference) were retrieved and used in sensitivity analyses. Exposure data were extracted to the level of precision reported in the published study before being harmonised to PA volume in mMET-hours/week. Risk estimates were extracted exactly as reported. Results reported separately by sex or other attributes  or for multiple cohorts within an article were treated as separate associations. If the original publication did not fully report information required for the meta-analysis, data were obtained from other publications using the same cohort, imputation procedures  or by coOSF repository. The last columns of the table on ‘Details of included articles and original and harmonized exposures’ in the same OSF repository contain the details of how the original PA exposure categories of each study were harmonised to mMET-hours/week values.A comprehensive data harmonisation was conducted to combine PA data from different sources. We harmonised reported PA exposure levels from all included studies into a common metric of non-occupational PA volume in mMET-hours/week, reflecting the rate of energy expenditure outside of work, above the resting metabolic rate (1 MET). This allows correct equating of PA volume of time spent at different PA intensity levels.PA exposures described in studies as frequency and duration measures were converted to weekly duration. If session duration was not provided, we assumed a duration of 0.75 hours/session . Categorical frequency data  were converted to weekly duration using assumptions for both frequency  and duration of sessions. When PA intensity was not explicitly described, we considered the reported activities as light, moderate or vigorous based on the description provided and using the Compendium of Physical Activities.For studies reporting energy expenditure without adjustment for body weight , energy expenditure was divided by reported weights to derive MET equivalents . If unavailable, body weight was calculated from reported BMI and height. If BMI was reported without height, a mean value from national survey data was used for height.To marginalise studies reporting volume of PA in gross units, 1 MET-hour was subtracted for each hour of activity. If the mean duration was not available, we used a conversion equation derived from all remaining studies where both volume and duration were available .Some articles provided aggregated exposures of non-occupational and occupational PA alongside other behaviours, such as sleep or sedentary time. When quantified information about these behaviours was available, this was subtracted from the point estimate of each exposure category. If no quantifiable data were available, we assumed occupational activity to be 40 hours/week at 1.25 MET (or 0.25 mMET) or the value of the lowest exposure category and subtracted this from all exposure categories.We conducted a meta-analysis for any outcome with results from at least four cohorts. Where necessary, risk estimates were recalculated to set the least active category as the referent.We performed a two-stage random-effects meta-analysis. In the first stage, we estimated study-specific dose–response coefficients using generalised least squares to incorporate the correlation within each set of log-relative risks.To investigate the potential effect of study-level confounders on the pooled results across outcomes, we conducted subgroup analyses with the 11 studies that reported results for all all-cause, CVD and cancer mortality outcomes and contrasted them with results from the corresponding main analysis. We also conducted subgroup analysis by sex using studies that reported separate results for men and women.PIFs were calculated for all outcomes based on PA exposure levels in the population of all included cohorts for a given outcome , using a fixed PA level of 8.75 mMET-hours/week, relevant to the six sources of bias.dosresmeta package (V.2.0.1).https://github.com/meta-analyses/.Analyses were performed using R V.4.0.5 and the Patients and the public were not involved in the development of this work.OSF repository, and data underlying each of the dose–response estimations can be downloaded at https://shiny.mrc-epid.cam.ac.uk/meta-analyses-physical-activity/.We screened titles and abstracts of 48 525 articles, of which 1280 were selected for full-text screening. A total of 196 articlesMost study participants reported non-occupational PA levels below 17.5 mMET-hours/week (76% of person-years), with almost all data below 35 mMET-hours/week (94% of person-years). https://shiny.mrc-epid.cam.ac.uk/meta-analyses-physical-activity/.Inverse, curvilinear dose–response associations between PA and most outcomes were observed, with stronger associations at lower volumes of PA. In most cases, diminishing marginal differences in risk and increasing uncertainty were observed at high PA volumes, particularly beyond 17.5 mMET-hours/week. Interactive dose–response curves and the exposure distributions are available at The association was stronger and more curvilinear for all-cause and CVD mortality than for cancer mortality , table 1A strong curvilinear association was observed for total CVD incidence (27% lower risk (95% CI 21 to 31) at 8.75 mMET-hours/week). However, the associations were weaker and more linear for the incidence of specific CVD outcomes , with the strongest association observed for coronary heart disease (21% lower risk (95% CI 16 to 26) at 8.75 mMET-hours/week) , table 1The association was weaker and more linear for total cancer incidence (12% lower risk (95% CI 8 to 15) at 8.75 mMET-hours/week). For the incidence of site-specific cancers, curvilinear and stronger associations were observed for head and neck, myeloid leukaemia, myeloma, and gastric cardia (from 35% to 22% lower risk at 8.75 mMET-hours/week). Weaker and more linear associations were observed for lung, liver, endometrial, colon and breast (from 16% to 5% lower risk at 8.75 mMET-hours/week). Non-significant associations were observed for bladder, oesophageal, prostate and rectal cancer , table 1Assuming a causal relationship between non-occupational PA and the examined health outcomes, shifting the distribution of non-occupational PA in the cohorts towards the equivalent recommended level of moderate-to-vigorous aerobic activity (8.75 mMET-hours/week) would have averted a substantial fraction of the population disease burden. For example, if all individuals accumulated at least 8.75 mMET-hours/week, then 15.7% (95% CI 13.2 to 18.2), 12.3% (95% CI 9.4 to 15.2) and 7.1% (95% CI 5.1 to 9.0) of all-cause, CVD-related and cancer-related deaths, respectively, potentially would have been averted. Additionally, 10.9% (95% CI 8.4 to 13.3) and 5.2% (95% CI 3.6 to 6.9) of all incident cases of CVD and cancer would have been prevented. Notably, 10.1% (95% CI 8.3 to 11.9) of all deaths would have been prevented if all adults achieved at least half this PA level .The pooled analysis from the 11 studies that reported results for all-cause, CVD and cancer mortality showed weaker associations for all-cause mortality and stronger associations for CVD and cancer mortality .Sex-stratified results were available for all-cause, CVD and cancer mortality and incidence of total CVD, coronary heart disease, stroke, total cancer, and cancers of the colon and rectum. There was some evidence of stronger associations in women than in men for all-cause and CVD mortality and CVD incidence, but the reverse for cancer mortality and incidence .Across the five outcomes with the largest number of separate results , 83%–93% of the studies used statistical adjustments to control for other morbidities at baseline, often combined with exclusion of adults with more serious pre-existing conditions. Studies that predominantly used statistical adjustments tended to report stronger associations for CVD mortality compared with studies that excluded adults with pre-existing conditions at baseline. Between 54% and 59% of the studies did not exclude mortality and disease events occurring within the first few years of follow-up, and those that did tended to report weaker associations for all-cause mortality and CVD mortality and incidence. The median follow-up time was 10 years for each outcome, with no evidence of bias between studies with longer and shorter follow-up time. Data imputation was required for 58%–64% of the studies, but no evidence of bias was found. The most frequently used procedures for exposure harmonisation were measurement unit conversions (42%–49% of studies) and the assumption of standard values for the intensity and duration of PA reported (39%–50%), but no evidence of bias was found. In 75% of the studies, the original exposure assessment did not include occupational PA and no evidence of bias was observed for studies requiring domain separation .In studies that did not report the duration or intensity of PA, assuming (where appropriate) shorter session duration (0.5 hours instead of 0.75 hours) and lower intensity (1 mMET lower for moderate and vigorous PA) did not materially alter the dose–response associations for most outcomes, especially analyses which included the largest number of separate results . When shSensitivity analyses using analytical models that did not include adjustment for adiposity showed stronger associations for all-cause and CVD mortality, and the incidence of total CVD, heart failure, and site-specific cancers of the endometrium, kidney, lung and rectum . NotablyWe present comprehensive dose–response meta-analyses of the associations between non-occupational PA and a wide range of mortality and disease outcomes. Our extensive exposure data harmonisation allowed inclusion of a larger evidence base than previous meta-analyses for 17 of 22 health outcomes investigated. For instance, our meta-analysis for all-cause mortality includes over 160 million person-years and 810 000 deaths. These are 17 and 7 times larger, respectively, than what were included in the previously largest dose–response analysis to date (9.39 million person-years and 117 000 deaths).We showed inverse non-linear dose–response associations for all-cause mortality and a range of CVD and cancer outcomes. This suggests that the greatest potential benefits can be achieved through small increases in non-occupational PA for those with an inactive lifestyle, with incrementally smaller additional benefits up to approximately 17.5 mMET-hours/week. Above this level, the evidence base was weaker. Shifting population levels of activity towards achieving the equivalent to 150 min/week of moderate-to-vigorous aerobic PA (8.75 mMET-hours/week) potentially would have prevented 16% of all premature deaths recorded.et al’set alet alCompared with previous studies, we found similar associations for all-causeCompared with the results from the GBD,It is surprising that the risk difference for all-cause mortality  is similar to that for total CVD (29%) and much larger than for total cancer mortality (15%). Equally, it is surprising that the association for total CVD incidence is stronger than that for coronary heart disease or stroke. Weaker associations would be expected for more composite outcomes  than for the outcomes within them for which a significant association has been established. Potential explanations include detection bias for some disease events, misclassification of causes of death , differential residual confounding, reverse causality by outcome, and different inclusion criteria by disease outcome, such as the inclusion of different disease groups which may have stronger associations with PA.Our risk of bias and sensitivity analyses showed that the methodological assumptions and decisions made for the pooled analyses did not significantly impact the dose–response associations observed. However, some methodological differences between the original studies seem to alter the magnitude of the pooled associations. Studies that did not exclude people with pre-existing conditions at baseline or mortality or disease events occurring within the first years of follow-up tended to observe stronger associations, as did studies with shorter follow-up, which may in part be indicative of reverse causality.A strength of our study was the use of comprehensive exposure data harmonisation methods that enabled the inclusion of a larger evidence base. This also allowed us to study health associations over a wider range of the PA exposure and increased the density of data in the area of the dose–response curves with high non-linearity. Another strength is the use of more sophisticated meta-analytical methods that allow the shape of the dose–response association to vary across the exposure range, rather than linear models of transformed exposure or grouping doses within high versus low PA or multiple exposure intervals. We improved the placement of spline knots compared with our previous workOur study also has limitations. There are levels of the PA exposure for which health associations are not reported in many of the original studies and hence there is still substantial uncertainty at those levels in the meta-analyses for some outcomes. Many of the included studies relied on self-reported questionnaires without validation or calibration data that could be used in the exposure harmonisation. In the exposure data harmonisation procedures, it was also sometimes necessary to make assumptions about parameters of PA, such as intensity and duration, where these were not explicitly reported. Nonetheless, our sensitivity analyses demonstrated the robustness of our results to the assumptions made. Measurement error is likely to lead to an underestimation of the true association between PA and the various outcomes, as demonstrated by the stronger associations observed in studies using device-based measures of PA.Our findings support the current PA recommendations of 150–300 min/week of moderate aerobic PA ,It should be noted that our exposure estimates are derived from a variety of self-reported questionnaires that capture mostly moderate and vigorous activities. These exposure estimates differ from those derived using device-based measures, which also record light-intensity and intermittent activities that are more difficult to recall.Future research should investigate the reasons for the apparent inconsistencies in dose–response associations between composite and individual disease endpoints. Although our risk of bias and sensitivity analyses showed robustness to the approaches we took during data handling and dose harmonisation, future studies could quantify methodological uncertainties  and propagate them in the aetiological analyses to provide more realistic uncertainty ranges for the dose–response associations. The evidence base was weaker for very high volumes of activity and further research is required to ascertain the shape of associations more reliably at the higher end of the PA continuum.We found evidence of dose-dependent associations between increasing non-occupational PA and a wide range of mortality, CVD and cancer outcomes. The strongest associations were observed for all-cause and CVD mortality, with weaker associations for the incidence of cancer, including variation by site. Appreciable population health benefits might be gained from increasing PA levels of people who are inactive to just half the current health recommendations, with further benefits for all reaching at least the recommended level and smaller additional benefits beyond that.Higher levels of physical activity are associated with lower rates of premature death and chronic disease outcomes.The shape of the dose–response association has been more difficult to determine and has not been established for a range of chronic diseases.Accurate estimation of the dose–response association between physical activity and disease outcomes is needed to assess the disease burden of physical inactivity and the potential impact of changes to population levels of physical activity.Our findings suggest an appreciably lower risk of mortality, cardiovascular diseases and cancers from the equivalent of 75 min/week or less of moderate-intensity aerobic physical activity .Our results include the first dose–response meta-analysis of nine site-specific cancers: bladder, esophageal, gastric cardia, head and neck, kidney, liver, lung, myeloid leukaemia and myeloma.One in 10 premature deaths could have been prevented if everyone achieved even half the recommended level of physical activity.


Furthermore, subgroup analysis revealed that the protective effect of nuts on NAFLD was more significant in females . In summary, our findings provide support for a protective relationship between nut intake and risk of NAFLD. Further exploration of the association between other dietary components and NAFLD is an important avenue for future research.Although previous epidemiological studies have been conducted to investigate the relationship between nut consumption and the risk of nonalcoholic fatty liver disease (NAFLD), the evidence remains inconclusive and contentious. The aim of our study was to further conduct a meta-analysis of observational studies to explore the latest evidence of the influence of nut consumption on NAFLD. This meta-analysis included a comprehensive search of all articles published in the PubMed and Web of Science online databases as of April 2023. A total of 11 articles were included, comprising 2 prospective cohort studies, 3 cross-sectional studies, and 7 case–control studies, and a random effects model was used to evaluate the relationship between nuts and NAFLD. Results showed that the odds ratio (OR) of NAFLD was 0.90 (95% CI: 0.81–0.99,  Nonalcoholic fatty liver disease (NAFLD) is a significant public health concern and is the primary cause of chronic liver disease. NAFLD is a liver manifestation of metabolic syndrome and is prevalent in obese and diabetic patients . With thMetabolic-associated fatty liver disease (MAFLD) is a novel concept proposed in 2020. Nut consumption is a crucial part of the Mediterranean diet and an important contributor of plant white matter, unsaturated fat acid (UFA), dietary fibers (DF), antioxidants and polyphenols, protein, vitamin E, and some trace elements (such as zinc and selenium) ,6. Nuts p = 0.0004) [Nut consumption has a protective effect on cognitive function, particularly among individuals at high risk of cognitive decline . Naghshi 0.0004) .Another cross-sectional study revealed that moderate intake of nuts and seeds significantly reduced the incidence of NAFLD in males (9%) and females (15%) compared with nonconsumers . Recent Previous meta-analysis has explored the relationship between nut consumption and the risk of various diseases, such as diabetes, obesity, and cancer ,27,28. HThe design and reporting in our study followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) . This woTo identify observational studies on the relationship between nut intake and NAFLD, we conducted a systematic search of the PubMed and Web of Science databases for articles published online prior to April 2023. Metabolic-dysfunction-associated fatty liver disease (MAFLD) is a novel concept proposed in 2020 . We alsoThe criteria for inclusion in this study were as follows: (1) observational epidemiological studies, cohort studies, case–control studies, or cross-sectional studies; (2) the correlation between nut intake and the risk of NAFLD; (3) data such as hazard ratio (HR), risk ratio (RR), or odds ratio (OR), and a reported confidence interval (CI) of 95%; (4) the intake of nuts  as exposure variables and the risk of NAFLD as the outcome variable. If more than 1 dataset was published in different articles, we selected the latest version for analysis. When referring to articles, we excluded the following: (1) letters, reviews, case reports, and comments; (2) studies that did not contain nut intake; (3) studies related to animals and cells; (4) studies that did not contain HR, RR, or OR ,31.L.P. and Y.X., independently screened the title, abstract, and full text of the study, retaining articles that met the criteria. Any differences were resolved through discussion, mediation, and consultation with a third researcher (J.S.). From each study, the year of publication, the name of the first author, the type of study, the duration of the study, the location of the study, the number of participants, their gender, confounding variables, and the OR, HR, or RR between nut intake and NAFLD risk, as well as the corresponding 95% CI, were extracted.Quality was assessed using the Newcastle-Ottawa Scale (NOS) scored on a scale of 0 to 9 . The scoI2. Acceptable heterogeneity of the results was defined as I2 ≤ 50%, whereas high heterogeneity was defined as I2 > 50%. When significant heterogeneity between studies was detected, subgroup analysis was conducted to identify possible sources of heterogeneity based on participant gender, sample size, geographic location, and study design. Publication bias was evaluated using Begg’s funnel plot and the Egger test. All statistical analyses were performed with Stata statistical software version 15.0.This study used OR or RR or HR to represent the magnitude of the effect. The random effects model was used to determine the heterogeneity of the articles by calculating Cochran’s Q and As shown in The characteristics included in the study are shown in p < 0.001) between nut intake and total NAFLD risk, indicating a negative correlation. Heterogeneity was observed in the study . Subgroup analysis explained the heterogeneity between studies by evaluating study design, gender, sample size, and nut intake. The asymmetric Egger linear regression test obtained a value of p = 0.463, and the Begg’s test was also used to obtain a value of p = 0.443. Therefore, there was no publication bias in this study.The study used a random effects model to calculate the relationship between nut intake and total NAFLD risk. The meta-analysis results are shown in p < 0.001). There was significant negative correlation between nut intake and NAFLD risk for women  when compared with the risk for men . In the subgroup analysis by region, we did not observe a correlation between nut intake and NAFLD risk in the Asian , Europe  and North American groups . In terms of study design, we observed a significant negative correlation between nut consumption and NAFLD in the cohort study  and cross-sectional studies . However, no significant correlation was observed in case–control studies . Finally, we also conducted a subgroup analysis based on the number of participants, with 1000 people as the grouping markers. Compared to the group with <1000 participants , the group with ≥1000 participants  had a significantly stronger correlation between the protective effect of consuming nuts on NAFLD.This study also conducted a subgroup analysis based on gender, region, and study design, and the results are shown in Nuts are rich in fiber, monounsaturated fatty acids (MUFA), and polyunsaturated fatty acids (PUFA), which could effectively prevent cardiovascular diseases and chronic diseases such as diabetes ,38. Recen-6: n-3 PUFA diet could upturn the metabolic phenotype of adolescents with fatty liver [n-6 PUFA can reduce liver fat and moderately improve metabolic status without losing weight [We have attempted to propose some mechanisms to explain the relationship between nut intake and NAFLD. The first mechanism involved antioxidant activity and inhibition of oxidative stress reactions. Nuts are rich in antioxidants and unsaturated fats , such asty liver . A case–g weight . Polypheg weight ,50,51. Ag weight . The inhg weight .p < 0.05) [In addition, consuming nuts can affect the risk of NAFLD through its effect on weight control. NAFLD is the most common liver manifestation of metabolic syndrome, closely related to obesity and insulin resistance ,55. Weig < 0.05) . Nuts ar < 0.05) . Tindall < 0.05) . Multipl < 0.05) ,61,62. T < 0.05) .Finally, the anti-inflammatory effect of nuts is also an important mechanism for achieving protective effects on a fatty liver. NAFLD is a general term that encompasses nonalcoholic fatty liver (NAFL) to nonalcoholic steatohepatitis (NASH) . Inflammp < 0.001). However, a study by Han et al. reported that nuts do not have a protective effect towards NAFLD in both males and females  [Our final study showed that consuming nuts can effectively reduce the risk of NAFLD  . The res50–3.39) ,25. Nuts50–3.39) . The exc50–3.39) ,72. This50–3.39) . Chronic50–3.39) .Our meta-analysis has several strengths. First, it includes a large sample size of 92,621 participants and provides a comprehensive analysis of the effect of edible nuts on NAFLD. Previous meta-analyses on nuts have primarily focused on their relationship with cancer incidence and mortality rates. Second, we conducted a stratified analysis to investigate the impact of various confounding factors on the relationship between nut consumption and NAFLD risk. In addition, we attempted to propose two potential mechanisms to explain the protective effect of nuts on NAFLD. To the best of our knowledge, our study is the first meta-analysis to investigate the relationship between nut consumption and the risk of NAFLD, which is a significant contribution to the field.However, when interpreting our findings, we must also consider its potential limitations. First, the number of studies analyzing the relationship between nuts and NAFLD was limited, with only 12 studies meeting our inclusion criteria. Due to the differences in study design, population characteristics, and other factors, there was a high degree of heterogeneity among the studies. In addition, the limited number of studies included in subgroup analyses may have led to biased results. Finally, as participants’ daily intake of food was not limited to nuts, but also contained other vegetables, fruits, and so on, it is possible that the potential impact of other dietary factors on the risk of NAFLD was not fully controlled for in the included studies, despite attempts to adjust for confounding variables. However, the protective effect of specific nuts on the liver could not be investigated because there were few studies. This will also be the focus of our future studies.Our findings suggested a significant association between nut intake and a reduced risk of NAFLD, indicating a protective effect. Although gender-specific data were limited, our findings suggested that nuts may have a stronger protective effect towards NAFLD in women. Despite the growing evidence of the health benefits of nuts, research on the relationship between nuts and NAFLD remains limited. To our knowledge, our study is the first meta-analysis to investigate the impact of nut consumption on NAFLD risk, and we proposed two potential mechanisms to explain the protective effect. Nuts could be an important dietary component in the prevention of NAFLD, and further research is needed to better understand the protective effects of nuts on NAFLD and the relationship between other food types and NAFLD.


Gastric cancer is the fifth most prevalent cancer in the world. The relationship between gastric cancer and underweight, overweight, and obesity is not fully understood yet. Obesity is a modifiable factor that has a positive association with some cancers. This systematic review and meta-analysis aimed to investigate the association between gastric cancer risk and abnormal body mass index, as an indicator of abnormal weight. Based on our results, obesity and overweight increase the risk of gastric cancer, while underweight is not associated. These findings can help policymakers and healthcare providers to make preventive strategies for controlling obesity and overweight.2: 89.08%) in underweight class, 1.155  in overweight class, and in 1.218  obesity class. There is no difference between cardia and non-cardia gastric cancer, while non-Asian race and female gender have higher risk of cancer, as Meta-regression of obesity and overweight classes showed. These findings suggest that there is a positive association between excess body weight and the risk of GC, with a higher impact in women than men and in non-Asian than Asian populations. Since abnormal weight is tied to various diseases, including GC, healthcare experts, and policymakers should continue interventions aiming to achieve a normal BMI range.This study aimed to investigate the risk of gastric cancer (GC) in abnormal body mass index (BMI) groups. A systematic search was carried out on Embase, PubMed/Medline, and Scopus from January 2000 to January 2023. The pooled risk ratio (RR) with a 95% confidence interval (CI) was assessed using a random-effect model. Thirteen studies with total of 14,020,031 participants were included in this systematic review. The pooled RR of GC was 1.124 (95% CI, 0.968–1.304, I Gastric cancer (GC) is the fifth most prevalent cancer in the world and accounts for one million one hundred thousand new cases and seven hundred sixty-nine thousand deaths in 2020 . AlthougOver the past three decades, obesity rates have risen progressively, with a three-fold increase reported by the World Health Organization (WHO) since 1975 . A recenDuring recent years, several systematic reviews have assessed the relationship between GC and BMI though they have yielded variable results ,29,30,31This study was carried out based on Preferred Reporting Items for Systematic reviews and Meta-Analyses , the PRIWe conducted a systematic search in three databases including Embase, Pubmed/Medline, and Scopus for studies reporting the association between GC and BMI from 1 January 2000 to 1 January 2023. We selected original studies with cohort methodology written in English language from initial pool by following MeSH terms: ‘Stomach neoplasms’, ‘Body Weight’, ‘Body Mass Index’, ‘Obesity’, ‘Lifestyle’, ‘Demography’, ‘Social Factors’, ‘Socioeconomic Factors’, ‘Sociodemographic factors’, ‘Metabolic Syndrome’, and ‘Risk Factors’ . In addiAll papers retrieved were imported into EndNote X8 , and studies meeting the following criteria were included: I. Cohort studies published between January 2000 to January 2023. II. Patients diagnosed with gastric adenocarcinoma (cardia or non-cardia). III. Patients aged over 20 years. IV. Papers providing data on GC incidence in relation to patients’ BMI based on either the WHO  or the Asian-Pacific category  in the main manuscript or Two blinded reviewers independently extracted data from eligible studies and entered it into a designed data extraction form. In case of discrepancies between the two reviewers, a third reviewer extracted the data. This form included label, first author, publication year, study location, study design, baseline characteristics (age and gender), type and number of the outcome, baseline BMI, follow-up time, definition of exposed and unexposed population, and the number of exposed and unexposed population.Two independent reviewers assessed the quality of the included studies using the Newcastle–Ottawa scale . The sca2 statistic. I2 values of more than 50% were considered high heterogeneity [p value < 0.05 was considered indicative of statistically significant publication bias and funnel plot asymmetry also suggested bias) [Statistical analyses were performed using Comprehensive Meta-Analysis software, version 3.7 . The pooled risk ratios (RRs) with 95% CI were assessed. The random-effects model was used because of the estimated heterogeneity of the true effect sizes. The between-study heterogeneity was assessed by Cochran’s Q test and the Ied bias) . A total of fourteen million twenty thousand and thirty-one participants were included in the analysis drawn from thirteen studies ,63,64,652: 95.18%) and 1.218 . Therefore, overweight and obesity could mildly increase the risk of GC. However, this number was insignificant in the underweight class . As between-study heterogeneity was high, we conducted a one-leave-out sensitivity analysis, which revealed that exclusion of any individual study did not significantly affect the overall findings, except for the study by Lim et al., in which exclusion resulted in a significant association between BMI and GC risk among underweight participants  and within-studies (gender) analyses to assess the interacting effects of demographics. The within-studies analysis is the preferred method to compare influential factors , and it Several meta-analyses investigated the effects of obesity and GC, but the results have been controversial ,30,31,32The risk of GC associated with obesity and overweight is twice as high in non-Asian subgroup compared to all studies (Obesity: 57% and overweight: 43%). A similar but lower increase was evident in two meta-analyses of prospective cohorts ,28. YangOne possible explanation for the lower role of BMI in GC risk among Asians is that other risk factors, such as Helicobacter pylori infection ,68, geneThe impact of screening programs on the relationship between BMI and GC in Asian and non-Asian populations is another possible factor that may contribute to the observed difference. Several Asian countries have implemented screening programs due to the high prevalence of gastric cancer in these areas . These pInterestingly, we found that females with excess weight have a higher risk of GC than males. In contrast, two systematic reviews in 2013 and 2014 found that males with obesity are more susceptible to GC than females by performing a “stratified meta-analysis” ,29. The The GC incidence itself shows a sex-difference, indicating the role of gender-related factors in carcinogenesis . On the The characteristics of GC differ between cardia and non-cardia GC. However, it is currently unclear whether this difference also applies to the relationship between obesity and GC as we could not confirm it. Two meta-analyses of case-control and cohort studies found that overweight or obesity was associated with an increased risk of cardia GC ,83 or noThe leading risk factor for cardia type is the high level of acid secretion in GERD, followed by gastritis caused by H. pylori infection ,85. Non-This study, similar to some other reviews, discovered carrying extra weight can lead to cancer ,28,29. SWe found that being underweight is not associated with GC risk, which is also evident in other cohort studies ,62. HoweOther explanations for the effect of BMI on the carcinogenesis of GC have been proposed. The span that a person’s BMI exceeds the normal range in life has a strong relationship with the risk of GC ,64,89. IOne should be aware of the limitations that need to be addressed before interpreting the results. I. The high level of heterogeneity among studies, which requires cautious interpretation of the results since future studies may change the direction of effect. The heterogeneity may be explained in part by variations in different gender and ethnic populations. II. Several potential confounding factors, including physical activity, dietary intake, H. pylori infection, and screening programs, could influence the results of the included observational studies. The presence of screening programs in Asian countries with high GC prevalence may alter the association between GC and other factors such as BMI. This could introduce bias since many patients benefit from early lesion removal and thus may not be included in the GC group. A meta-analysis of individual patient data may be warranted in the future since a meta-analysis itself cannot adjust for the effect of confounding factors. III. All non-English results and precancerous outcomes were excluded, which can cause bias. IV. Different BMI classifications caused the exclusion of some studies from the meta-analysis and subgroup analysis. V. All included cohorts are conducted in developed countries in Northern America, Europa, and East Asia. Therefore, results should be used in similar settings and contexts, while studies from developing countries are recommended. VI. The time of GCs’ onset after starting overweight or obesity could not be assessed, which may be of great importance.According to this study, obesity and overweight can increase the risk of GC, whereas there is no significant relationship between underweight and GC. The impact of excess body weight on GC risk is higher in women and non-Asian populations. However, high heterogeneity is evident in the association between abnormal weight and GC risk, which may be explained in part by variations in different gender and race populations. Some unexplained questions should be addressed in future basic and epidemiological research. Since abnormal weight is tied to various diseases, including GC, healthcare experts and policymakers should continue interventions aimed at achieving a normal BMI range.