Screening adherence is important in reducing colorectal cancer (CRC) incidence and mortality. Disparity in CRC screening adherence was observed in populations of different socioeconomic status (SES), but the direction and strength of the association remained unclear.We aimed to systematically review all the observational studies that have analyzed the association between SES and adherence to organized CRC screening based on fecal occult blood tests.We systematically reviewed the studies in PubMed, Embase, and Web of Science and reference lists of relevant reviews from the inception of the database up until June 7, 2023. Individual SES, neighborhood SES, and small-area SES were included, while any SES aggregated by geographic areas larger than neighbors were excluded. Studies assessing SES with any index or score combining indicators of income, education, deprivation, poverty, occupation, employment, marital status, cohabitation, and others were included. A random effect model meta-analysis was carried out for pooled odds ratios (ORs) and relative risks for adherence related to SES.Overall, 10 studies, with a total of 3,542,379 participants and an overall adherence rate of 64.9%, were included. Compared with low SES, high SES was associated with higher adherence . In the subgroup of nonindividual-level SES, the adjusted association was significant . However, the adjusted association was insignificant in the subgroup of individual-level SES . As for subgroups of the year of print, not only was the unadjusted association significantly stronger in the subgroup of early studies  than in the subgroup of late studies , but also the adjusted one was significantly stronger in the early group  than in the late group , which was consistent and robust. Despite being statistically insignificant, the strength of the association seemed lower in studies that did not adjust for race and ethnicity  than the overall estimate .The higher-SES population had higher adherence to fecal occult blood test–based organized CRC screening. Neighborhood SES, or small-area SES, was more competent than individual SES to be used to assess the association between SES and adherence. The disparity in adherence between the high SES and the low SES narrowed along with the development of interventions and the improvement of organized programs. Race and ethnicity were probably important confounding factors for the association. Worldwide, colorectal cancer (CRC) ranks third in cancer incidence and second in cancer mortality . It has The benefits of screening depend highly on screening compliance . What isA review conducted by de Klerk et al  qualitatThere is no conclusive evidence on the association between SES and adherence in FOBT-based organized screening. It is warranted to examine the association between SES and adherence, contributing to the improvement of screening adherence and equity to promote health equity in screening programs. Thus, the study aimed to perform a meta-analysis focusing on the association between SES and FOBT-based organized CRC screening adherence.We conducted the systematic review and meta-analysis under the direction of the PRISMA  statement . We searFirst, titles and abstracts, and then full-text articles, were screened for inclusion by 2 investigators (ZL and CW) independently. We included studies that measure the association between SES and adherence to organized CRC screening programs. Inclusion and exclusion criteria are shown in Inclusion criteriaOriginal articlesObservational studies conducted in organized colorectal cancer (CRC) screening programsUsed fecal occult blood tests (FOBT) as the only primary screening testAverage-risk populationAssessed socioeconomic status (SES), including individual SES, neighborhood SES, and small-area SESAssessed SES with any index or score combining indicators of income, education, deprivation, poverty, occupation, employment, marital status, cohabitation, and others, and the SES index or score was defined as categorical variablesProvided data to obtain related either unadjusted effect sizes (including odds ratios and relative risks) or adjusted onesAssessed adherence or nonadherence to FOBT-based CRC screening, which was objectively recordedAssessed adherence rate with the percentage of individuals invited to participate in screening that participateEnglish languageExclusion criteriaLacked enough data to calculate either unadjusted or adjusted effect sizes, including qualitative studies, reviews, commentaries, letters, and guidelinesConference abstracts that did not contain enough information for quality assessmentAssessed adherence to colonoscopy or other screening testsFocused on only low-SES population or high-SES population, for example, focusing on only low-income populationIf several studies were conducted in the same program, participants should be counted only once. Duplicated samples were excluded, and the study with the biggest sample that offered the most information was included.Assessed country-level SES, province-level SES, or any SES aggregated by geographic areas larger than neighborsAssessed self-reported adherenceAny language other than EnglishSample data were extracted by 2 investigators (ZL and CW) independently, with disagreements solved by discussion and consensus. The standardized form consists of (1) the first author; (2) publication year; (3) study design; (4) country; (5) program name; (6) sample size; (7) adherence rate; (8) name of the SES index or score; (9) SES measures involved; (10) focusing on SES or not; (11) SES level; (12) original adjusted effect size; (13) available unadjusted and adjusted effect sizes (relative risks were transformed into odds ratios [ORs]) ; (14) coThe Newcastle-Ottawa Quality Assessment Scale (NOS) for cohort studies and the Agency for Healthcare Research and Quality (AHRQ) methodology checklist were separately used to assess the quality of cohort and cross-sectional studies, which categorized the studies as high, middle, and low-quality -28. The Q statistics and I2 estimations were used to assess study heterogeneity. With considerable heterogeneity, a random effect model using the Daimonian and Laird method was used to conduct the meta-analysis.ORs with corresponding 95% CIs were used for analysis, and an inverse variance method was used. We used Stata  to analyze data. The “flocci” command was used to convert the reference group of original estimates. “mean,” “meta bias,” and “meaning” were used to perform analysis. The Cochrane We first assessed the unadjusted bivariate correlations between SES and adherence. Then we used the effect sizes of multivariate models with the most adjustment factors to assess the adjusted association. Tables and forest plots were used to display the results.Q statistic and the I2 statistic. Subgroup analysis was conducted to probe the potential source of heterogeneity across studies according to some characteristics of the studies: year of print (before or after the median year of print of the studies), adherence rate , SES foidelines ,30. SensThe initial search yielded 1819 articles, and 9 articles were identified from reference lists. Of the 1819 articles, 910 remained after removing 909 duplicates by EndNote X8.1 (Thompson Reuters) automatically  24,31].,31.24,31A total of 9 studies provided unadjusted effect sizes, with 1,927,075 (63.7%) participants in 3,026,991 targeted individuals. And 8 studies reported adjusted effect sizes, involving 3,220,822 targeted individuals and 2,117,839 (65.8%) participants.A positive and strong association was observed between SES and adherence . The forest plots are shown in The positive association between SES and adherence remained significant in all feasible subgroup analyses (with at least three studies in a group), except for the adjusted association in the subgroup of individual-level SES  than in late studies  seemed smaller than the overall pooled estimate . The adjustment for race and ethnicity better showed a positive association. In addition, the strength of the association between SES and adherence seemed lower when the adherence rate was low or when SES was the focus of the studies  than not .To our knowledge, this is the first systematic review and meta-analysis assessing the strength of the association between SES and FOBT-based organized CRC screening adherence. It indicated considerably higher adherence among the highest SES compared with the lowest. The positive association was significant in both unadjusted and adjusted analyses, while being consistent across most possible sets of pooled estimates and all possible subgroup analyses, including the year of print, adherence rate, SES focus, SES level, and adjustment for race and ethnicity.The association between high SES and high adherence was similar to those reported in 4 previous reviews investigating the participation of multiple modalities of CRC screening in both organized and opportunistic screening programs -45. FurtSubgroup analysis of different SES levels indicated that individual-level SES was not associated with adherence, which suggested that different SES levels could be the source of heterogeneity between studies. A systematic review by Mosquera et al  indicateNevertheless, country-level SES, which was not involved in this meta-analysis, was confirmed to not be associated with adherence either. In 2009, Pruitt et al  conducteDue to the difficulty and trouble of collecting individual-level data, commonly used measures of SES were based on geography, at different levels of aggregation from countries to neighborhoods . And it Although socioeconomic and racial and ethnic disparities in organized CRC screening were noted ,51, the The results of this meta-analysis showed that the disparity between the high and low-SES populations in adherence decreased. Since the adherence rate of the subgroup with a later year of print was higher than that of the other subgroup, a probable explanation was that adherence of the low-SES population increased more than the high as organizational screening programs developed with more and more interventions to promote adherence. Actually, a large-sample randomized controlled trial proved that certain interventions could not only improve overall participation but also particularly enhance the adherence of the low-SES population . In thisThis is the first meta-analysis assessing the association between SES and adherence, especially with the first focus on FOBT-based organized CRC screening adherence. Previous reviews merely focused on the qualitative association between SES and adherence and did not deal with the influence of confounding factors. However, this study confirmed, quantified, and further explored the positive association between SES and adherence by performing meta-analysis and subgroup analysis and comprehensively interpreting the results of both the unadjusted association and the adjusted one.A few limitations should be considered when interpreting the results. First, considerable heterogeneity existed and although we performed subgroup analysis and sensitivity analysis, it remained. The probable explanation of the heterogeneity was that in the absence of a widely recognized SES index, we observed different kinds of indexes and scores, which combined different indicators at different levels and used different weights to indicate SES. However, most indexes and scores incorporate the 3 SES indicators of income, education, and employment. The different start ages and stop ages used by the different programs resulted in differences in the mean age of the population, leading to differences in sample characteristics. Second, we recognized that all the studies included were carried out in the high-income setting even though we did not make a restriction on the inclusion criteria and actively supplement the literature collection from the reference lists. It might be caused by a limited number of studies that quantitatively assessed the association between SES and adherence to organized screening programs, especially in low-income countries where organized screening was not widespread. Therefore, whether the results can be generalized to low- and middle-income countries remains unknown. Finally, different adjustments might cause residual confounders in the included studies. To minimize this effect, we took both unadjusted and adjusted effect sizes into account and were more confident to draw a conclusion when the results of them were consistent.To conclude, we found the high-SES population had higher adherence to FOBT-based organized CRC screening. Neighborhood SES or small-area SES was more competent than individual- and country-level SES to be used to assess the association between SES and adherence. Race and ethnicity were probably important confounding factors for the association. The good news was that the disparity of adherence between the high SES and the low SES narrowed with the development of interventions and the improvement of organized programs. Future research should focus on digging into the causation of the association and creating targeted interventions and strategies to improve the adherence of the low-SES population, which aims at overcoming the inequality in the chances of benefiting from organized screening programs.


P = .03) and the MR-Egger analysis showed an odds ratio of 5.81E-11 . We found no significant heterogeneity or pleiotropy, and sensitivity analysis confirmed the robustness of our results. Furthermore, statistical power calculations suggested that our findings were reliable. Conclusion The use of PPIs is a protective factor for pancreatic malignancies, but no causal relationship has been found with other digestive system tumors.The objective of this study was to explore the causal relationship between the use of proton pump inhibitors (PPIs) and 16 types of digestive system tumors. We utilized a 2-sample Mendelian randomization (MR) approach to investigate this relationship. We obtained exposure and outcome data from the UK Biobank and the Finland Biobank, respectively. The genetic data used in the analysis were derived from genome-wide association studies (GWAS) studies conducted on European populations. We screened single nucleotide polymorphisms significantly associated with the use of omeprazole, a commonly used PPIs, as instrumental variables. We then performed MR analyses using the inverse variance weighting (IVW) method, MR-Egger regression, and the weighted median method to evaluate the causal effect of omeprazole use on the 16 types of digestive system tumors. Our MR analysis revealed a significant causal relationship between the use of omeprazole and pancreatic malignancies, but not with any other types of digestive system tumors. The IVW analysis showed an odds ratio of 4.33E-05 (95%CI: ,  Proton pump inhibitors (PPIs) are a commonly used treatment for acute and chronic acid-related diseases of the digestive system, including gastroesophageal reflux disease, Zollinger-Ellison syndrome, peptic ulcer, upper gastrointestinal bleeding, and Helicobacter pylori infection. They work quickly and effectively to suppress gastric acid secretion. However, the potential adverse consequences of frequent and long-term PPIs use should be taken seriously, as they are widely used in clinical practice. The question of whether PPIs use increases the risk of developing digestive system tumors is a topic of debate within the international community. In contrast, a Korean population-based cohort study found no association between PPIs use and an increased risk of gastric cancer. A meta-analysis suggested that the use of PPIs may be associated with an increased risk of pancreatic malignancies. On the other hand, another meta-analysis found no conclusive evidence of a link between PPIs use and pancreatic malignancies. A significant nested case-control study conducted in the United States found no association between exposure to proton pump inhibitors for more than 2 years and an increased risk of gastrointestinal tumors. In contrast, results from another comprehensive analysis suggested an association between PPIs use and a higher risk of gastric, pancreatic, and liver cancer, but not esophageal or colorectal cancer.A population-based study showed that long-term PPIs users had a 2.4-fold increased risk of gastric cancer compared with non-users. This increased risk was not observed among histamine 2-receptor antagonist users.Overall, the multiple observational studies and meta-analyses mentioned above have failed to reach a clear consensus, possibly due to differences in regions, populations, baseline characteristics, and follow-up times of various cohorts, which may confound the results and conclusions of these studies. Furthermore, most studies only demonstrate a correlation and do not establish a significant causal relationship. Therefore, the current evidence is insufficient to support concerns about the carcinogenic side effects of PPIs. This study employs 2-sample Mendelian randomization to examine the potential causal relationship between PPIs use and digestive system tumors, with the goal of providing a reference for future clinical medication.Mendelian randomization (MR) is a technique that utilizes genetic variations as instrumental variables to mitigate potential biases that may arise in observational clinical studies and account for confounding factors. This method is frequently employed to infer causal relationships between exposure and outcomes. Data from 2020 substantiates its prevalent use, with prescription counts surpassing 56 million. This volume situates Omeprazole as the eighth most frequently prescribed pharmaceutical in the United States.,10 For this study, omeprazole was selected as the exposure factor, with benign and malignant tumors of the esophagus, stomach, small intestine, colon, rectum, anus and anal canal, liver and bile duct, and pancreas serving as the outcomes. Two-sample Mendelian randomization was utilized to analyze all data, using publicly available summary statistics from genome-wide association studies (GWAS).In the 23rd edition of the World Health Organization Model List of Essential Medicines, published in 2023, Omeprazole is distinguished as the sole PPIs included.To satisfy the 3 fundamental assumptions of 2-sample Mendelian randomization  with the data field “20003_1140865634,” while data for 16 types of digestive system tumors were obtained from the FinnGen Biobank in the IEU open GWAS database (https://gwas.mrcieu.ac.uk/datasets/?gwas_id__icontains=finn-b). To avoid potential confounding effects from other cancers, all digestive system tumors were excluded from other cancer types. Refer to Table To mitigate potential biases based on ethnicity and gender, exposure and outcome data were obtained from GWAS data of European male and female populations. To reduce the risk of sample overlap, omeprazole data were obtained from the UK Biobank  were selected as instrumental variables based on their association with the exposure. Due to the constraint of the number of SNPs, those with ,15 in R software (Version 4.2.1), with a significance level of α = 0.05.To analyze the data in this study, 3 Mendelian randomization methods were utilized: the weighted median estimator, inverse variance weighting (IVW), and MR-Egger regression. The IVW method was used as the primary analysis method, while weighted median estimator and MR-Egger were employed as secondary methods. All analyses were conducted using the TwoSampleMR 0.5.6 packageP > .05 indicates the absence of heterogeneity. For multiplicity analysis, the intercept of the MR-Egger regression was evaluated using the mr_pleiotropy_test function. An intercept that tends to 0 and P > .05 suggest the absence of horizontal pleiotropy. Sensitivity analysis was conducted using the leave-one-out method, which involved excluding 1 SNP at a time and repeating the IVW analysis.To assess heterogeneity, the Cochran Q statistic was calculated using the IVW and MR-Egger methods, and the results were depicted on a funnel plot.  to validate the reliability of the results.The statistical power of the MR analysis was assessed using the online tool mRNDP = .03) and a MR Egger . In other words, the use of omeprazole might be a protective factor for pancreatic malignancies, whereas there was no significant causal relationship with the other 15 digestive system tumors. Additionally, a scatter plot  and the MR-Egger method  did not reveal any significant heterogeneity in the heterogeneity test. A funnel plot  and the MR Egger method . The results were robust, as evidenced by the heterogeneity test, pleiotropy analysis, sensitivity analysis, and statistical power calculation. In summary, taking omeprazole can reduce the incidence of pancreatic malignancies, and there was no significant causal relationship with the remaining 15 digestive system tumors.This study utilized the Mendelian randomization method to investigate the causal relationship between omeprazole and the incidence of 16 digestive system tumors. The results indicated that omeprazole use was a protective factor against pancreatic malignancies, as supported by the IVW method , was unavailable for detailed subgroup analysis. Finally, the causal relationship between omeprazole and pancreatic malignancies identified in this study is only theoretical and does not elucidate its biological mechanism.In summary, previous studies have often regarded PPIs use as a risk factor for digestive system tumors. This study employed 2-sample Mendelian randomization methods with genetic variation as an instrumental variable to investigate the causal relationship between omeprazole use and the incidence of 16 digestive system tumors. The study found evidence that omeprazole use could reduce the risk of pancreatic malignancies incidence, but no causal relationship was detected between omeprazole use and esophageal or stomach tumors, among other types of digestive system tumors. Therefore, the evidence for the long-term use of PPIs increasing the risk of developing digestive system tumors remains inconclusive, and additional research is needed to explore its potential mechanisms for preventing tumor occurrence. This study has significant clinical implications for the safe use of PPIs and can provide new insights for future clinical and cell-based research. Larger-scale prospective cohort studies, randomized controlled experiments, or cell experiments are necessary to generalize and apply these findings.Data curation: Ruiqi Zhao, Mengyao Han, Zhimei Lin.Funding acquisition: Lisheng Peng.Investigation: Zhimei Lin, Mengjiao Yu.Methodology: Ruiqi Zhao, Sen Lin, Lisheng Peng.Project administration: Lisheng Peng.Resources: Ruiqi Zhao, Mengyao Han, Mengjiao Yu.Software: Ruiqi Zhao, Mengyao Han.Supervision: Sen Lin, Mengyao Han, Zhimei Lin, Mengjiao Yu, Lisheng Peng.Validation: Sen Lin, Mengyao Han.Visualization: Ruiqi Zhao.Writing – original draft: Ruiqi Zhao.Writing – review & editing: Ruiqi Zhao, Sen Lin, Lisheng Peng.


Higher MD adherence was significantly associated with a reduced GC risk in male , but not in female . Furthermore, adherence to the MD possibly decreased the risk of gastric cardia adenocarcinoma (GCA)  and gastric non-cardia adenocarcinoma (GNCA) . Our results indicate that adherence to the MD reduces the risk of GC and its subtypes.Available results on the association between the Mediterranean diet (MD) and gastric cancer (GC) incidence are controversial. The present study aimed to determine the correlation between different subtypes of GC and MD adherence. This meta-analysis was registered on PROSPERO (CRD42021284432). We searched Embase, PubMed, Cochrane Library, and Web of Science from inception through 22 April 2023 to retrieve relevant studies. A random-effects model was used to pool odds ratios (ORs) with 95% confidence intervals (CIs). Eleven studies were included in the meta-analysis. Pooled analyses revealed that adherence to the MD was inversely associated with GC risk (OR Gastric cancer (GC) is a serious health concern worldwide. According to the International Agency for Research on Cancer’s data on the latest global burden of cancer, new GC cases in the world reached over one million in 2020, ranking fifth among the overall new cancer cases . GC is aDiet and multiple chronic non-communicable diseases are strongly linked ,4. IncreFurthermore, GC can be categorized into gastric non-cardia adenocarcinoma (GNCA) and gastric cardia adenocarcinoma (GCA) according to the anatomical site . The twoThis meta-analysis complied with the Preferred Reporting Items of Systematic Reviews and Meta-Analyses (PRISMA) guidelines . It was We retrieved all relevant articles published before 22 April 2023, from Embase, Cochrane Library, PubMed, and Web of Science. The specific search strategy is shown in The following studies were eligible for our study: (1) cohort or case–control studies; (2) studies reporting the association between the MD and the risk of GC; (3) studies on any type of GC; and (4) studies with adjustment of hazard ratios (HRs), relative risks (RRs), or odds ratios (ORs) with corresponding 95% confidence intervals (CIs). If several publications reported similar clinical findings, the one with the most complete data was included in the present analysis.The exclusion criteria were as follows: (1) non-English publications; (2) unavailability of the full text; and (3) review articles, conference abstracts, letters, case reports, systematic reviews, and meta-analyses.Two researchers (X.B. and X.L.) independently checked the titles and abstracts of the searched studies and excluded apparently irrelevant articles. Any disagreement was settled by discussion. The following data were extracted from the included studies: author’s name, study design, year of publication, country, population, follow-up (cohort studies), number of cases and controls (case–control studies), age, gender, components of score, adjustment and risk estimates, and In this meta-analysis, two researchers (X.B. and X.L.) used the Newcastle–Ottawa Scale  to assescc) and cohort studies (ORcoh) were also represented, respectively. And the estimates of observational studies were derived from the according estimates of both case–control and cohort studies. As is known, cohort studies often have a follow-up time compared to case–control studies, and previous studies have shown a lower effect estimates with increasing follow-up time [Stata 16.0  was used for statistical analysis. A DerSimonian and Laird random-effects model was used to pool multivariate-adjusted ORs with 95% CIs ,15. In t-up time . Additio2 statistics, with I2 values of 25%, 50%, and 75% graded as low, medium, and high heterogeneity, respectively [p < 0.05 indicated statistically significant differences.Heterogeneity was measured using the chi-squared test (Cochran’s Q test) and Iectively . To furtectively . The folectively  and Eggeectively . A two-sDuring the initial search, 1987 relevant studies were identified. After removing duplicates, the titles and abstracts were checked. Next, 120 full-text articles were evaluated for eligibility. We finally included eleven studies (with five case–control studies and six cohort studies) in this meta-analysis after reviewing the full text ,27,28,29cc, 0.43; 95% CI, 0.29 to 0.63, I2 = 85.2%; ORcoh, 0.84; 95% CI, 0.77 to 0.92, I2 = 10.4%; ORobservational, 0.68; 95% CI, 0.57 to 0.81, I2 = 82.9%) [A random-effects model was used for data analysis, and the results showed that high MD adherence was significantly correlated with a reduced risk of GC (OR= 82.9%) ,27,28,29Data from R2 = 0.53, p = 0.001, which is highly significant. Therefore, it is not reasonable to pool case–control and cohort studies without considering the effect of follow-up time on the effect estimates, and instead we chose to take the results of the according case–control studies as the primary results for MD and the risk of GC and its subtypes.The linear regression is OR = 0.427 + 0.028 × years, adjusted cc, 0.64; 95% CI, 0.49 to 0.83, I2 = 86.1%) and GNCA . However, there was no risk reduction for GCA in cohort studies  [GC was divided into two subtypes based on tumor sites: GCA and GNCA. The subgroup analysis revealed that the highest MD adherence was correlated with a lower risk of GCA (OR= 65.7%) ,23,27,28Subgroup analysis was performed to further explore the source of heterogeneity. Neither geographic location nor gender changed the direction of effect estimates, and the results for the source of heterogeneity are presented in cc, 0.21; 95% CI, 0.11 to 0.39, I2 = 0.0%) [cc, 0.52; 95% CI, 0.36 to 0.76, I2 = 87.3%) [Based on the geographic location of the included studies, the subgroup analysis showed that adherence to the MD significantly reduced GC risk in Asia (OR = 0.0%) ,25. This= 87.3%) ,29  [cc, 0.83; 95% CI, 0.68 to 1.01, I2 = 57.1%) [The subgroup analysis also showed that the MD had a protective effect on GC in males OR= 75.5%) . However5.5% . Ho= 57.1%)  . This finding indicated that geographic location and gender does explain the source of heterogeneity.Subgroup analyses showed that the Asia and female subgroups had lower heterogeneity  (p = 0.108)  A. There = 0.108) B.As far as we know, diet plays a key role in GC development, which is gradually attracting the attention of both oncologists and nutritionists in the field of cancer. Numerous meta-analyses have investigated the correlation between a specific food and GC risk ,31. GeneIn addition to particular items in the diet, GC risk is influenced by dietary patterns. The MD pattern is characterized by simple cooking and light flavors, with more emphasis on whole grains, vegetables, seafood, and fruits, and less emphasis on red meat and sweets, as recommended by nutritionists. Schwingshackl et al.  reportedIn this meta-analysis, the inverse association between MD and GC risk can be found in observational studies, including case–control and cohort studies. However, a recent review on vitamin D and cancer stated that observational studies using 25-hydroxyvitamin D concentrations drawn prior to cancer diagnosis are believed to be more precise than blood drawn closer to the time of cancer diagnosis . In otheTo the best of our knowledge, Helicobacter pylori infection is a major risk factor for GNCA; however, no positive correlation between H. pylori infection and GCA has been reported ,43. A prParticipants’ diet data were collected from the included studies using a food frequency questionnaire . The MD Other limitations also affected the interpretation of the results. First, the adjustment factors in the included studies were inconsistent, leading to a large difference in the value of OR and HR. This affected our statistical results. Second, all eligible studies were prospective cohort or case–control studies without explicit intervention, and the number of included studies was too small. Third, MD may differ in each country, which also resulted in high heterogeneity. Studies included in this meta-analysis used different MD scales and diet pattern analyses, causing it to not be possible to compare results and draw a causality among different scales. However, our results showed a high relationship between adherence to different types of MD scales and GC risk. We hope that future studies would minimize these biases and verify our results. Moreover, additional randomized controlled studies with larger sample sizes and better design are needed to investigate causality.The present meta-analysis demonstrates that high adherence to MD is correlated with a lower GC risk. Additionally, higher MD adherence is associated with a decreased risk of both GCA and GNCA. Future well-designed studies are required to confirm our results.


Idiopathic normal-pressure hydrocephalus (iNPH) is a treatable cause of dementia; however, its etiology and pathogenesis remain poorly understood. The objective of this study was to investigate the prevalence and impact of vascular risk factors in patients with iNPH compared to a control cohort to better understand the potential mechanisms and preventive measures.We systematically searched PubMed, Web of Science, Embase, and the Cochrane Library  for studies reporting vascular risk factors for the development of iNPH. Pooled odds ratios (ORs) and 95% confidence intervals (CIs) were estimated using random-effects models.N = 991, OR = 2.30, 95% CI 1.64 to 3.23, I2= 64.0%), diabetes mellitus (DM) , coronary heart disease , and peripheral vascular disease  increased the risk for iNPH, while overweight was a possible factor  based on the sensitivity analysis. Smoking and alcohol consumption were not associated with iNPH.After screening 1,462 articles, 11 case-control studies comprising 1,048 patients with iNPH and 79,668 cognitively unimpaired controls were included in the meta-analysis. Our data showed that hypertension (Our study suggested that hypertension, DM, CHD, peripheral vascular disease, and overweight were associated with iNPH. These factors might be involved in the pathophysiological mechanisms promoting iNPH. These findings require further investigation in future studies.https://www.crd.york.ac.uk/PROSPERO/, CRD42022383004. Idiopathic normal-pressure hydrocephalus (iNPH) is a clinical syndrome characterized by cognitive decline, gait disturbance, and urinary incontinence, with ventricular enlargement apparent on brain imaging . A recenSome observational studies have indicated that about one in four patients with iNPH have vascular risk factors . At presThis study was conducted in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines . We haveStudies that met the following criteria were included: (1) patients with iNPH with at least one of the features of the Hakim's triad  and radiologically confirmed ventricular enlargement according to the existing diagnostic criteria ; (2) stuStudies were excluded if they: (1) were reviews, meeting abstracts, editorial materials, or articles not published in English; (2) the odds ratios (ORs) and 95% confidence intervals (CIs) were unextractable; (3) included participants with symptomatic or secondary NPH, including cerebrovascular diseases, head trauma, brain tumors, or infections; and (4) only evaluated patients with asymptomatic ventricular enlargement.https://www.ohri.ca/programs/clinical_epidemiology/oxford.asp). A “star system” was devised to evaluate studies, with a focus on three aspects: the selection of the study groups, comparability of the groups, and ascertainment of either the exposure or outcome of interest in the context of case-control or cohort studies, respectively. Two investigators (HL. C and F. Y) independently assessed all eligible studies, and disagreements were resolved through consensus. Articles with 8–9 stars were rated as high quality, 5–7 stars as moderate quality, and 4 stars as low quality  was performed by eliminating one study at a time to evaluate the stability of the results and explain the possible sources of heterogeneity. Publication bias was assessed by visually inspecting the funnel plot and statistically examining the results using the Begg's and Egger's tests if more than five studies were synthesized for each factor. P < 0.05 was considered statistically significant. All statistical analyses were performed using Stata SE 16.0 .A pooled OR with a 95% CI was calculated for patients with iNPH based on possible vascular risk factors. The ogeneity . Since cIn total, 1,462 publications were retrieved after the initial search. The titles and abstracts of 1,043 studies were screened after removing duplicates, leaving 69 articles that were assessed for eligibility by reading the full text. Fifty-eight records were excluded for several reasons. Finally, 11 studies were included in the meta-analysis .Detailed characteristics of the included studies are presented in N = 991). The studies defined hypertension based on previous medical history, and in most (80%), the cut-off value was set at 140/90 mmHg, except for two early studies that used 160/95 mmHg and 160/90 mmHg   . Sensiti= 0.003) . The met= 0.003) .N = 985) reported a relationship between DM and iNPH development. DM was mostly defined (89%) based on medical history or medication use, with only one study using the results of a glucose tolerance test  (Nine studies (nce test . The med= 0.075) . Sensiti= 0.075) . Meta-re= 0.075) .N = 880) suggest that CHD may be a risk factor for the development of iNPH  . Pooled results from seven studies (= 0.000) . Sensiti= 0.000) .N = 364) suggest that smoking was not a risk factor for the development of iNPH  (The combined results of the six studies (= 0.422) . Sensiti= 0.422) .N = 225) discussed the relationship between overweight and the development of iNPH, and the results suggested that overweight was a risk factor for iNPH    . Two stu.440)  (2 (2 . Sensiti= 0.659) .N = 234) investigated the relationship between hyperlipidemia and iNPH and defined hyperlipidemia using different criteria. One study found that elevated Apolipoprotein B/A1 ratio was relevant with iNPH  (P = 0.37) or history of hyperlipidemia  and iNPH (Three studies N = 23 investig) P = 0.3 or histoand iNPH .N = 77) examined the relationship between alcohol use and iNPH development, and the combined results showed that alcohol consumption was not a risk factor for iNPH  (Only two studies ( = 0.01) .N = 172) investigated the relationship between self-reported peripheral vascular disease and iNPH development and found that it was associated with the iNPH development  (Two studies (= 0.861) .We visually inspected funnel plots and statistically used the Begg's and Egger's tests to evaluate publication bias for hypertension, DM, CHD, and smoking. As shown in funnel plots, no significant publication bias was detected .In this study, we comprehensively reviewed the vascular risk factors for iNPH and identified five modifiable factors associated with iNPH. Hypertension was the most common vascular comorbidity, followed by DM, CHD, and peripheral vascular disease. Although overweight was also considered a potential vascular risk factor, its association with iNPH remained inconclusive owing to unstable sensitivity analysis results. In contrast, based on a few studies, smoking, alcohol consumption, and hyperlipidemia were not associated with iNPH.The relationship between hypertension and iNPH has been reported in previous studies , 10, 12.Several factors may explain the association between DM and iNPH. In mouse models, DM has been linked to neuroinflammation, waste accumulation, and reduced aquaporin 4 (AQP4) density, resulting in impairment of the glymphatic system , 34. ThiOur meta-analysis suggested that overweight may be a risk factor for iNPH. Although no obvious heterogeneity was detected, instability in the results was found on omitting a large prospective case—control study . This maIn addition, we found that both CHD and peripheral vascular disease were risk factors for iNPH. However, heterogeneity may arise from variations in the diagnostic criteria across studies. Because these diseases are mainly atherosclerotic in nature, atherosclerosis may play a crucial role in the pathogenesis of iNPH . A previGiven the high prevalence of vascular comorbidities among individuals with iNPH, it becomes imperative for clinicians to understand the potential impact of these factors on the surgical outcome. Vascular risk factors, such as hypertension, diabetes mellitus, coronary heart disease, peripheral vascular disease, cerebrovascular disease, and smoking, tend to exert a detrimental influence on the prognosis of iNPH patients –54. HoweOur study had several limitations. First, most of the included studies were case-control studies, and establishing causality was challenging. Therefore, long-term follow-up studies are required. Second, some ORs were calculated from raw data without adjusting for potential confounders; we could not exclude unknown variables that affected the results. Third, publications written in languages other than English and conference proceedings were excluded, which might have resulted in a publication bias.Further research is required to better understand the etiology and pathogenesis of iNPH. First, the current study had a cross-sectional design; therefore, we could not explore a causal relationship between these vascular risk factors and the development of iNPH. In addition, it is unclear whether these modifiable risk factors affect the prognosis of patients with iNPH. Long-term follow-up studies are needed to clarify these issues. Second, novel modifiable vascular risk factors, such as chronic kidney disease, physical inactivity, obstructive sleep apnea, inflammatory markers, and cerebral small vessel disease, also need to be explored to better understand disease mechanisms. Third, an internationally unified algorithm for diagnosing iNPH is required to compare different studies. Finally, despite conducting a systematic search, most studies were conducted in Western countries, and the current research lacks data from low- and middle-income countries; therefore, epidemiological characteristics and differences among these regions are urgently needed.In this systematic review, we identified five modifiable vascular risk factors in patients with iNPH: Management of hypertension, DM, CHD, overweight, and peripheral vascular disease. These factors may be involved in the pathophysiological mechanisms promoting iNPH. Further studies are required to confirm these findings.The original contributions presented in the study are included in the article/HC and FY conceived this study and drafted the manuscript. HC, FY, HG, KH, LQ, and RW collected the information and relevant materials. QC designed the study and revised the manuscript. All coauthors revised the manuscript and approved the final version.


It is unclear what causes the increased incidence of non-Hodgkin lymphoma (NHL); however, chemical substance exposure is known to be one of the risk factors for the disease. The aim of our systematic review was to verify the association between occupational exposure to carcinogens and NHL risk. In our literature review, 51 articles were included in the meta-analysis resulting in an overall OR of 1.27 (95% CI 1.04–1.55). Among these studies, 20 reported a significant association with the increased risk of NHL. We demonstrate that the risk of NHL increases for individuals occupationally exposed to pesticides, benzene, and trichloroethylene. Our findings may provide information for public health and practical decision-making about certain work activities and the use of chemical compounds.Non-Hodgkin lymphoma (NHL) is a heterogeneous group with different types of diseases. It remains unclear as to what has led to an increase in incidences of NHL, however, chemical substance exposure is known to be one of the risk factors for the disease. Therefore, we performed a systematic review and meta-analysis including case-control, cohort, and cross-sectional observational epidemiological studies to verify the association between occupational exposure to carcinogens and NHL risk. Articles between the years 2000 and 2020 were collected. Two different reviewers performed a blind selection of the studies using the Rayyan QCRI web app. Post-completion, the selected articles were extracted and analyzed via the RedCap platform. Our review resulted in 2719 articles, of which 51 were included in the meta-analysis, resulting in an overall OR of 1.27 (95% CI 1.04–1.55). Furthermore, it was observed that the main occupation associated with the increased risk of NHL was that in which workers are exposed to pesticides. We therefore conclude that the evidence synthesis of the epidemiological literature supports an increased risk for NHL, regardless of subtype, considering occupational exposure to certain chemical compounds, mainly pesticides, benzene, and trichlorethylene, and certain classes of work, primarily in the field of agriculture. Non-Hodgkin lymphomas (NHL) are a heterogeneous group of diseases that stem from the proliferation of malignant lymphocytes and their precursors. These cells accumulate in the lymph nodes, but can also extend to other organs ,2. ThereEtiologies of most NHL’s remain largely unknown, however, risk factors are likely to include immune deficiencies, Epstein-Barr virus, HTLV1, and Helicobacter pylori bacteria, radiation, chemicals, such as benzene, persistent organic pollutants, and pesticides ,6. SeverMany recent studies have examined the association between environmental or occupational exposure to chemical compounds and the risk of developing NHL and specific subtypes ,9,10,11.There have been several review and meta-analysis studies focusing on the association between environmental pollutants and NHL ,20,21,22The systematic review and meta-analysis were performed according to the guidelines specified by PRISMA ,24. The Meta analysis was conducted to identify which types of workers were occupationally exposed to chemical agents that were potentially linked to the development of non-Hodgkin lymphomas. A systematic literature search of articles in the Medline, Cochrane Library, Virtual Health Library-BVS Regional Gateway, Scopus, PubMed, and Embase databases, from 2000 to 2020, was conducted. The search included the keywords “environmental carcinogens”, “iarc classification”, “occupation cancer”, “non-Hodgkin’s lymphoma”, “carcinogenic agents”, “environmental factors”, “occupational cancer”, “occupational exposure”, “workers”. Details of this search strategy are reported in Studies that fulfilled the specific inclusion criteria were included in the meta-analysis: (a) case-control, cross-sectional, prospective cohort, and retrospective type observational epidemiological studies; (b) original studies based on workers exposed to carcinogens; studies whose outcome was NHL; (c) articles written in English only; (d) studies in which the association between occupational exposure and the risk of NHL were expressed as relative risk (RR), standardized mortality ratios (SMR), hazard risk (HR), standardized incidence ratios (SIR), proportional mortality ratio (PMR), odds ratio (OR) with 95% confidence intervals (CIs), either reported or could be obtained from the data reported in the article. Studies such as meta-analysis, reviews, meetings, abstracts, letters, and comments were not included in this review. Study selection was performed independently by two reviewers using the blind system using the Rayyan QCRI online web app software  to categStudy quality assessment was carried out using the Study Quality Assessment Tool, developed by methodologists at the National Heart, Lung, and Blood Institute (NHLBI), in conjunction with professionals at the Research Triangle Institute International. The tool was based on other assessment tools created by the Evidence Based Practice Centers of the Agency for Healthcare Research and Quality (AHRQ), Cochrane, the Scottish Intercollegiate Guidelines Network and, others working in the context of evidence-based medicine. Designed to assist reviewers, the tool was considered, for this research, fundamental to the critical evaluation of the internal validity of a study. This includes items to assess potential flaws in study methods, including risks of bias.During the quality review of the studies, each item of the tool may be answered with a “yes”, “no”, or “cannot be determined/not reported/not applicable”. For each item in which “no” was selected, a potential risk of bias was considered. This assessment tool is not designed to provide a list of factors that make up a numerical score or to delimit a score for reviewers. Thus, it is used, by following with the literature, for intervention studies, a rating of bad ≤ 6, regular > 6 and 5, and good ≥ 10 is used. For case-control studies that contained 12 questions instead of 14, the rating was adjusted: bad ≤ 5, regular > 5, and <9, and good ≥ 9 .Results from the cohort, cross-sectional, and case-control studies that associated exposure to chemical compounds or work class with increased risk of NHL were included in the meta-analysis. The meta-analysis was performed for each comparison using the random effects model, and the OR with the 95% Confidence Interval (CI) was calculated using the Mantel Haenszel method. The combined OR (ssion of : (1)ψ^MHkw = (k ckb)/knK = total number of studies k = 1, 2,…, Kkψ = (ak dk)/(bk ck)kn = study sample size kka = number of events in the group exposed in the study kkb = number of non-events in the group exposed in the study kkc = number of events in the unexposed group in the study kkd = number of non-events in the non-exposed group in the study kkn = sample size 2) [Variables included in the meta-analysis were sample size, number of exposed individuals, number of non-exposed individuals, number of exposed individuals with an event, and non-exposed individuals with an event. Heterogeneity between each study was estimated using the Q-statistic of the Q-Cochran test and quantified using the inconsistency index (I2) . Funnel 2) . AnalyseThe search of the database resulted in 2719 articles. Duplicate and irrelevant articles were excluded as described in the inclusion criteria. A total of 537 duplicate articles were excluded while the remaining 2182 were selected for evaluation based on titles and abstracts. Evaluation of the titles and abstracts revealed that 1854 articles did not analyze the association between occupations and carcinogens with the increased risk of NHL. As a result, an additional 1854 articles were excluded. The remaining 328 articles were subjected to a full-text review; of these, 27 articles did not meet the inclusion criteria. During data extraction, 129 articles were excluded by methodological evaluation, as a result, only 172 remained eligible for the study. Ultimately, 51 articles were included in the meta-analysis, because 121 articles did not present the necessary information for the calculation of the effect size (Odd Ratio) .The systematic review comprised 76 case-control articles and 96 cohort/transversal articles. In regards to geographic coverage, of the 172 articles included in this review, 74 studies were conducted in the United States, 17 in Canada, 14 in France, 13 in Italy and Sweden, 10 in Germany and Spain, 9 in Denmark and Norway, 7 in China and the United Kingdom, 6 in Australia, New Zealand, and the Czech Republic, 5 in Finland and Ireland, 4 in Japan, 3 in Brazil, England, and Iceland, 2 in Korea, Greece, the Netherlands, and Iran, 1 in South Africa, Singapore, Poland, Taiwan, and Ukraine .Several studies conducted at these locations are cohort studies or are part of consortia. The Agricultural Health Study (AHS) was the main cohort study evaluated in the studies ,40,41,42The association between a large number of work classes and chemical compounds associated with the development of NHL and its subtypes has not been evaluated in any review. Reviews typically evaluate the association between the increased risk of overall NHL or some specific subtype with only one group of chemical agents or work class ,22,46,47A total of 51 observational studies that met our inclusion criteria were identified and included in the meta-analysis. Accordingly, 28 case-control studies and 23 cohort studies were included. p < 0.01, indicating significant heterogeneity across studies . This suggests that some work classes and occupational exposure to certain compounds are associated with a 27% increase in the risk of NHL. The highest OR entered was for the study by Mahajan et al.  OR = 14. The ran studies .Altogether, more than 20 different work classes and more than 30 chemical compounds were evaluated . Of the Although pesticides constitute the main exposure evaluated in the articles, four of the 20 articles that revealed significant OR had evaluated exposure to solvents, formaldehyde, trichloroethylene, and benzene, which were the main compounds analyzed in these articles. In addition, significant OR was also observed for studies that evaluated employees in the manufacture and operation of aircraft, employees exposed to meat, and factory workers exposed to benzene . FormaldGiven the high heterogeneity observed in our meta-analysis, we carried out subgroup analyses as an approach to identify the potential sources of heterogeneity in our overall meta-effects estimate. Thus, combined estimates of studies by NHL subtype, experimental design types, and exposure types were conducted. In the analysis performed for each NHL subtype evaluated in more than three studies, included in our meta-analysis, a meta-OR above 1.0 with significance for the random effects model was only observed for the DLBCL  .The most frequent subtype of non-Hodgkin lymphoma worldwide is DLBCL which presents as a heterogeneous disease group with variable outcomes . In our The remaining meta-analyses for the remaining subtypes generated meta-OR above the null value of 1.0, yet, with no significance . Some meA significant association between exposure to any single solvent and the risk of FL was also identified . In our When analyzing the different types of study designs, only the case-control studies analysis revealed a meta-OR above 1.0 with significance for the random-effects model , while for the meta-analysis performed with the cohort studies, the meta-OR was less than 1.0 and non-significant  . For theA meta-analysis, which included studies that analyzed only work class, did not exhibit significantly increased risk of NHL  . DespiteFor the meta-analysis that included studies that examined only chemical compounds, significantly increased risk of NHL was observed  . In thisIn addition to pesticide exposure, exposure to solvents, such as benzene (n = 1) , trichloThe meta-analysis that included studies that examined both work class and chemical compounds did exhibit a significantly increased risk of NHL  . The stuSubgroup analyses such as NHL subtype, study design, and exposure type did not reveal the leading sources of heterogeneity observed in our overall meta-analysis . NeverthOur meta-analysis synthesized 51 epidemiological studies examining the relationship between chemical agents and work classes and the risk of NHL. To the best of our knowledge, this is the first and largest systematic review with meta-analysis that takes into account all exposure types  and all subtypes of NHL. Overall, our results provide evidence for the hypothesis that occupational exposure to chemical agents increases the risk of NHL. Thus, these results represent an important contribution to the literature on exposures associated with the development of NHL.A limitation of the study is that our review was conducted with studies published between 2000 and 2020 . However, according to the new classification of the World Health Organization to Hematolymphoid Tumors: Lymphoid Neoplasms (5th edition), the term LNH is no longer used . The 51 In this systematic review through meta-analysis, we present the evidence, through a detailed evaluation of epidemiologic studies, supporting the association between occupational chemical exposure and the risk of developing non-Hodgkin lymphoma. We demonstrate that the risk of NHL, regardless of the subtype, increases for individuals occupationally exposed to chemical agents, mainly pesticides, benzene, and trichloroethylene, as well as for certain work classes, primarily for occupations in agriculture. However, there is still insufficient data on the association between NHL and specific chemical compounds. Our findings may provide information for public health and practical decision-making about certain work activities and the use of chemical compounds. Furthermore, the evidence for the association of specific chemical compound classes and work classes associated with the development of NHL in biological samples is still limited, so future mechanistic studies, measuring exposures, and evaluating the biological and molecular effects associated with the risk of NHL are still needed.


TERT (rs2736100), CCDC26 (rs4295627), CDKN2A/B (rs4977756) and RTEL1 (rs6010620) gene polymorphisms as the risk factors specific to glioma. However, the data samples of previous genetic-epidemiological studies are modest to determine whether they have definite association with glioma.Previous genetic-epidemiological studies considered The study paid attention to systematically searching databases of PubMed, Embase, Web of Science (WoS), Scopus, Cochrane Library and Google Scholars. Meta-analysis under 5 genetic models, namely recessive model (RM), over-dominant model (O-DM), allele model (AM), co-dominant model (C-DM) and dominant model (DM) was conducted for generating odds ratios (ORs) and 95% confidence intervals (CIs). That was accompanied by subgroup analyses according to various racial groups. The software STATA 17.0 MP was implemented in the study.TERT gene rs2736100 polymorphism, CCDC26 gene rs4295627 polymorphism, CDKN2A/B gene rs4977756 polymorphism and RTEL1 gene rs6010620 polymorphisms increased the risk of glioma in Caucasians to different degrees. In Asian populations, the CCDC26 gene rs4295627 polymorphism and CDKN2A/B gene rs4977756 polymorphism did not exhibit a relevance to the risk of glioma. It is suggested to cautiously explain these results as the sample size is small.21 articles were collected. According to data analysis results, in four genetic models  TERT (rs2736100), CCDC26 (rs4295627), CDKN2A/B (rs4977756) and RTEL1 (rs6010620) genes in glioma might increase risk of glioma, but there are ethnic differences. Further studies evaluating these polymorphisms and glioma risk are warranted.The current meta-analysis suggested that the SNP of  Based oTERT), 8q24.21 , 9p21.3 , and 20q13.33  (TERT) is crucial for telomerase activity in preserving telomeres and cellular immortality (CCDC26) downregulates telomerase activity and increases apoptosis. Cyclin-dependent kinase inhibitor 2A/B (CDKN2A/B) can partially control the cell cycle. Regulator of telomere elongation helicase 1 (RTEL1), as a DNA helicase, can inhibit homologous recombination for directly maintaining the genomic stability  . Telomerortality . In conttability . These gTERT (RS2736100), CCDC26 (RS4295627), CDKN2A/B (RS497756), and RTEL1 (rs6010620).In this meta-analysis, we expanded sample size by enrolling up 21 articles and used five genetic models to examine the relationship between glioma and SNPs of 2The meta-analysis in the study was performed as per the Preferred Reporting Items for Systematic reviews and Meta-Analysis version.2.1Literature search was carried out comprehensively in eight databases, namely PubMed, Embase, WoS, Scopus, Cochrane Library, and Google Scholars. The MeSH term were as follows: “ OR (SNP) OR  OR (SNP) OR (SNP) OR (SNPs) OR (variation) OR (mutation)) AND ((Gliomas) OR (Glioma) OR  OR  OR  OR  OR (Mixed Glioma) OR  OR  OR  OR (Astrocytoma) OR (Glioblastoma) OR (Diffuse Intrinsic Pontine Glioma) OR (Ependymoma) OR  OR (Ganglioglioma) OR (Medulloblastoma) OR (Oligodendroglioma) OR (Optic Nerve Glioma)) AND (rs xxxxxxx)”. In addition, we screened relevant reviews and references from previous meta-analyses to increase the number of other eligible studies. The retrieval process was performed by two researchers independently. Any discrepancy was adjudicated by a senior investigator.2.2TERT gene, rs4295627 polymorphism of CCDC26 gene, rs4977756 polymorphism of CDKN2A/B gene and rs6010620 polymorphism of RTEL gene with glioma susceptibility;(1) case-control or cohort study and genome-related study regarding the association of rs2736100 polymorphism of (2) studies containing allelic and genotypic data or ORs and 95% CIs for susceptibility;(3) appropriate statistical methods and reliable data with clear and unambiguous expression of results, allowing calculation of the ORs and 95% CIs;(4) no overlapped data, and studies involving the same data or overlapped by the same authors, only the one that had the largest sample size or the most recently published.2.3(1) review, commentary, abstract, and case report types;TERT (rs2736100), CCDC26 (rs4295627), CDKN2A/B (rs4977756) and RTEL1 (rs6010620);(2) articles with original literature content did not involve (3) articles were unrelated to glioma research;(4) articles with insufficient data, such as those lacking gene frequency information or ORs with 95% CIs;(5) articles with non-human study subjects.2.4Each study covered the data of: the first author’s name, publication year, ethnicity , number of cases and controls with relevant genotypes, ORs and 95% CIs. In addition, the genetic model used in the article was checked. We applied the Newcastle-Ottawa Scale for evaluating the literature quality. Two researchers performed independent data extraction and evaluated the literature quality. A senior investigator took charge of discrepancy arbitration.2.5p<0.05 reported statistical significance). Second, we measured the association degree of TERT (rs2736100), CCDC26 (rs4295627), TERT (rs2736100) and RTEL1 (rs6010620) with glioma risk under five genetic models using the ORs and 95% CIs, which were the RM, O-DM, AM, C-DM and DM.STATA 17.0 MP was implemented for all statistical analyses in this study. First, we performed the Hardy-Weinberg Equilibrium (HWE) test .We merged the ORs and 95% CIs that satisfied the criteria under the same models. In addition, in all genetic models, subgroup analyses were carried out according to ethnic differences and the 33.1We initially searched 640 literature and eventually enrolled 21 eligible articles covering Caucasian, Asian, and others , 17–36. 3.23.2.1TERT gene rs2736100 polymorphism, allele G is the risk gene (TERT gene rs2736100 polymorphism led to increased glioma risk in 5 genetic models (AM (G vs T): OR=1.29, 95% CI: 1.25 to 1.34; RM (GG vs TT+TG): OR=1.31, 95% CI: 1.18 to 1.44; DM (TG+GG vs TT): OR=1.52, 95% CI: 1.34 to 1.73; C-DM (GG vs TT): OR=1.69, 95% CI: 1.56 to 1.82; C-DM (GT vs TT): OR=1.40, 95% CI: 1.31 to 1.49; O-DM (TT+GG vs TG): OR=0.93, 95% CI: 0.88 to 0.98). When stratifying the analysis according to ethnicity, we obtained similar results in Caucasians, although in the O-DM results only showed a trend of association with glioma (AM (G vs T): OR=1.28, 95% CI: 1.23 to 1.33; RM (GG vs TT+TG): OR=1.29, 95%CI: 1.13 to 1.47; DM (TG+GG vs TT): OR=1.51, 95% CI: 1.29 to 1.76; C-DM (GG vs TT): OR=1.64, 95% CI: 1.50 to 1.80; C-DM (GT vs TT): OR=1.37, 95% CI: 1.27 to 1.49; O-DM (TT+GG vs TG): OR=0.93, 95% CI: 0.87 to 1.00). Among Asians, we included only one study, and the results were similar to those of Caucasians. In the study of a multi-ethnic population, we included the study of Wrensch and Safaeian which revealed the relevance of the TERT gene rs2736100 polymorphism to increased glioma risk.In the isk gene Figure 23.2.2CCDC26 gene rs4295627 polymorphism, allele T was the risk gene : OR=0.92, 95% CI: 0.81 to 1.03. In one study, the RM (GG vs TT+TG): 0.52; 0.25~1.05; DM (TG+GG vs TT): 0.86; 0.60 to 1.23; C-DM (GG vs TT): 0.50; 0.24 to 1.05; C-DM (GT vs TT): 0.95; 0.65 to 1.38; O-DM (TT + GG vs TG): 0.97; 0.67 to 1.40.3.2.3CDKN2A/B gene rs4977756 polymorphism, allele T was the risk gene : OR=1.03, 95% CI: 0.91 to 1.17. In one study, the RM (GG vs AA+AG): 1.12; 0.53 to 2.34; DM (AG+GG vs AA): 0.96; 0.66 to 1.39; C-DM (GG vs AA): 1.09; 0.52 to 2.32; C-DM (GA vs AA): 0.94; 0.63 to 1.38; O-DM (AA+GG vs AG): 1.08; 0.73 to 1.58. In other study, RM (GG vs AA+AG): 0.68; 0.25 to 1.81; DM (AG+GG vs AA): 0.98; 0.60 to 1.58; C-DM (GG vs AA): 0.69; 0.25 to 1.87; C-DM (GA vs AA): 1.04; 0.63 to 1.73; O-DM (AA+GG vs AG): 0.93; 0.56 to 1.52.3.2.4RTEL1 gene rs6010620 polymorphism, allele T was the risk gene : OR=1.35, 95% CI: 1.22 to 1.48; RM (GG vs AA+AG): 2.04; 1.61 to 2.57; DM (AG+GG vs AA): 1.35; 1.20 to 1.52; C-DM (GG vs AA): 2.41; 1.09 to 3.04; C-DM (GA vs AA): 1.22; 1.07 to 1.38; O-DM (AA+GG vs AG): 0.93; 0.83 to 1.05.3.3Our study adopted Egger’s Test for evaluating the literature publication bias and the results are shown in In rs2376100, there was publication bias in AM (p=0.469). In rs4295627, publication bias existed in dominant, co-dominant and O-DM (DM (TG+GG vs TT): p=0.426; C-DM (GT vs TT): p=0.436; O-DM (TT+GG vs TG): p=0.427). In rs4977756, publication bias existed in O-DM (p=0.111). In rs6010620, publication bias existed in the recessive, dominant and O-DM (RM (GG vs AA+AG): p=0.010; DM (AG+GG vs AA): p=0.014; O-DM (AA+GG vs AG): p= 0.461).4Our study is one of the first ones with conducting the first meta-analysis with the largest sample size as far as we concern and we performed detailed analyses of multiple associated SNPs under five genetic models. Besides, stratified analyses were carried out considering ethnicity. This indicates reliable meta-analysis results.From the meta-analysis, allele G in the SNP at the rs2736100 locus differentially elevated glioma risk, which conforms to studies by Di Stefano . In partTERT is the most important component in regulating telomerase activity. A few scholars found the relevance of TERT expression to glioma grade and patients’ prognosis  presented shorter telomeres relative to people who possessed TG genotype  and p15 (CDKN2B) oncoproteins which play multiple roles in cellular stress recognition, senescence regulation, differentiation as well as apoptosis in the developmental and proliferative phases of cells  regions of genes. Intron SNPs possibly facilitate aberrant exon inclusion or deletion by altering mRNA splicing and the formation and addition regarding nonsense transcripts (RTEL1 gene rs6010620 on 20q13.33 is a vital candidate genetic variant that has been widely reported (RTEL1 was a direct target of miR-4530. The abnormal expression of RTEL1 could lead to obvious reversion of the miR-4530 overexpression in glioma cell lines. The miR-4530/RTEL1 axis acts as a latent treating target specific to gliomas (RTEL1 has been linked to the development of astrocytoma and Hoyeraal-Hreidarsson syndrome (RTEL1 can maintain genomic stability by studying its ability to maintain telomere homeostasis and promote DNA replication, there are still a lot of questions that shall be highlighted. Nowadays, researches have not confirmed the regulation of RTEL1 and its recruitment to impact the replication forks and telomeres (nscripts , 57. Thereported . As repo gliomas . Not onlStudying the association between the risk of glioma and SNPs is definitive, because it could help clinicians use such polymorphisms as pragmatic molecular biomarkers of glioma risk, assist drug developers innovate relevant targeted therapeutic agents for the benefit of patients, and allow for basic researchers to explore the molecular mechanisms of glioma pathogenesis and accurately regulate glioma progression. According to meta-analysis results, racial differences exist in the risk of glioma, but there is a paucity of studies on Asian ethnicity, which requires researchers to conduct relevant studies. Exploration of SNPs at other loci and risk of glioma is also necessary.5Some limitations should be noted when interpreting the current results. The study also presents some limitations, which shall be noted during the interpretation of current studies.First, there were scarce studies on Asian populations. Studies on results in Asian populations should be interpreted with caution.In addition, the absence of more specific individual information about interaction within genes and that between gene and environment prohibited us from conducting a more precise analysis.Finally, some unavoidable publication bias in the results of meta-analysis should also be taken into account.6TERT gene rs2736100 polymorphism, CCDC26 gene rs4295627 polymorphism, CDKN2A/B gene rs4977756 polymorphism and RTEL1 gene rs6010620 polymorphism might all increase the risk of glioma development, but there are ethnic differences. It is necessary to well develop case-control studies with large sample sizes and a focus on more races or glioma types to overcome the limitations described earlier to make the conclusions more accurate.In summary, our study suggested that the All authors contributed to the article and approved the submitted version. YQW: Writing articles, searching literature and extracting literature. JZho and ZT: Analyzing the quality of articles. JZha: Analyzing Data. XC: Providing methodology. SL: Editing of articles.


Several risk factors of immune checkpoint inhibitors (ICIs)-associated acute kidney injury (AKI) have been reported sporadically. To identify the risk factors of ICIs-associated AKI in a large-scale population, therefore we conducted a systematic review and a real-world retrospective study.We search literature concerning risk factors of ICIs-associated AKI in ClinicalTrials.gov and electronic databases  up to January 2022. Meta-analysis was performed by using odds ratios (ORs) with 95%CIs. In a separate retrospective pharmacovigilance study by extracting data from US FDA Adverse Event Reporting System (FAERS) database, disproportionality was analyzed using the reporting odds ratio (ROR).A total of 9 studies (5927 patients) were included in the meta-analysis. The following factors were associated with increased risk of ICIs-associated AKI, including proton pump inhibitors(PPIs) , angiotensin-converting enzyme inhibitors (ACEIs)/ angiotensin receptor blockers (ARBs) , nonsteroidal anti-inflammatory drugs (NSAIDs) , diuretics , diabetes mellitus , genitourinary cancer , combination therapy of ICIs  and extrarenal immune-related adverse events(irAEs) . Furthermore, analysis from FAERS database verified that concurrent exposures of PPIs , ACEIs/ARBs , NSAIDs  or diuretics  were observed significant signals associated with AKI in ICIs-treated patients.Concurrent exposures of PPIs, ACEIs/ARBs, NSAIDs or diuretics, diabetes mellitus, genitourinary cancer, combination therapy, and extrarenal irAEs seem to increase the risk of AKI in ICIs-treated patients.The online version contains supplementary material available at 10.1186/s12882-023-03171-9. Strategies with immune checkpoint inhibitors (ICIs) to treat malignancy have been adopted with a significant improvement in patients' prognosis in recent years. However, ICIs might induce a series of immune-related adverse events(irAEs) due to the unrestricted activation of the immune system and the off-target mode of these drugs . Among tSeveral risk factors of ICIs-associated AKI have been reported recently, including the concurrent drugs of proton pump inhibitors (PPIs), angiotensin-converting enzyme inhibitors (ACEIs)/ angiotensin receptor blockers (ARBs) or diuretics, combination therapy with different ICIs, use of ipilimumab or pembrolizumab, coexisting chronic kidney disease (CKD) or low estimated glomerular filtration rate(eGFR), hypertension, the combination of other irAEs and so on –5, 9–14.To identify the risk factors of ICIs-associated AKI in a large-scale population, therefore we conducted a systematic review of observational studies and a real-world study by extracting data on concurrent drugs from the US Food and Drug Administration (FDA) Adverse Event Reporting System (FAERS) database.Firstly, a meta-analysis of observational studies was conducted to investigate the risk factors of AKI in patients treated with ICIs. Secondly, a real-world analysis was performed by extracting data from the FAERS database to further verified the risk of concurrent drugs of ICI-associated AKI in a large-scale population.This study was registered prospectively in PROSPERO (CRD42021293326). The meta-analysis was performed according to the Preferred Reporting Items for Systematic Reviews and Meta-analysis (PRISMA) statement .The search for literature was conducted in the following databases: PubMed, Embase, the Cochrane Library, and ClinicalTrials.gov. We performed a systematic literature review of observational studies about ICIs-associated AKI (up to 10th January 2022). The terms searched for literature were that combined “immune checkpoint inhibitor” or “ipilimumab” or “tremelimumab” or “pembrolizumab” or “nivolumab” or “cemiplimab” or “atezolizumab” or “durvalumab” or “avelumab” with “acute kidney injury” or “AKI” or “acute renal failure” or “renal failure acute”.The inclusion criteria were as follows: 1) observational studies that compared clinical data of patients with versus without AKI after ICIs administration. The definitions and severity of AKI in each included study were demonstrated in supplementary table After duplicate records were removed, all abstracts were screened by two independent reviewers (P.C and J.Z), and potential eligible articles were searched for full texts. These two review authors also independently extracted data from each study that fulfilled all the criteria. Discrepancies were resolved by a third investigator (M.F). The following information was collected: 1) Sample size; 2) Baseline characteristics: median age with interquartile range (IQR) or mean age with standard deviation (SD), gender, race, baseline eGFR; 3) ICI regimens; 4) Cancer types ; 5) Comorbidities such as diabetes, hypertension, congestive heart failure, chronic obstructive pulmonary disease(COPD), liver disease; 6) Concurrent medications of PPIs, ACEIs/ARBs, nonsteroidal anti-inflammatory drugs(NSAIDs) and diuretics. Quality assessment of each included study was simultaneously performed by two authors(P.C and J.Z) with the Newcastle–Ottawa Scale (NOS) . Discrep2 statistic was used to present between-study heterogeneity, which of > 50% was considered substantial heterogeneity. Random-effect model was utilized when significant heterogeneity existed. Statistical analyses were performed with STATA software (version 16.0).Risk factors associated with ICIs-associated AKI were assessed with odds ratio (OR) with 95% confidence interval (CI). The IThe real-world pharmacovigilance study was based on the US FAERS database, which defines adverse drug reactions using Preferred Terms from the Medical Dictionary for Regulatory Activities . We only used the MedDRA Preferred Term “Acute kidney injury” to identify AKI cases as previous studies –19. More025) exceeded 1 and with at least 3 reports of AKI simultaneously. Statistical analyses were performed with Microsoft Excel .Disproportionality analysis was applied to assess whether AKI was differentially reported in ICIs-treated patients with or without target concurrent drugs. Reporting odds ratio (ROR) was utilized as reported previously , 20. RORAs demonstrated in Fig. Detailed characteristics of the included studies were summarized in Table Of these 9 included studies, 5 studies attained a score of 8 with NOS method, 2 studies of 7, 1 study of 6, and 1 study of 5 , ACEIs/ARBs , NSAIDs  and diuretics  were observed significant signals associated with AKI Table .Table 4DUntil now, the present study seems to be the first study to identify risk factors of ICIs-associated AKI by a systematic review of clinical studies and analysis of data from a worldwide pharmacovigilance database. Our study suggested that concurrent drugs exposure , coexisting diabetes mellitus, genitourinary cancers, combination therapy of ICIs, and extrarenal irAEs were associated with increased risk of AKI events in ICIs-treated cancer patients.PPIs have been reported to increase the risk of AKI in the general population with an abundance of evidence . DepositACEIs and ARBs both have effects on vasodilation of the renal efferent arterioles thus causing a reduction of glomerular filtration pressure. During the hypovolemia state, the reduced efferent vascular tone as described above may induce AKI . NeverthNSAIDs were reported to be correlated with an increased risk of AKI in the general population in both children and adults. The possible reasons were as follows: NSAIDs can reduce renal blood flow; and may cause tubular obstruction owing to crystal deposition then induce direct cytotoxicity and cell-mediated immune attack , 26. AmoDiuretics are reported to be a risk factor of AKI after liver transplantation and related with a higher risk of perioperative AKI , 28. DiuCoexisting diabetes mellitus DM is a risk factor of AKI in a certain status, such as infection with coronavirus disease-19, liver transplantation, percutaneous coronary intervention, and so on –31. HoweAKI was reported to occur in genitourinary cancers with a relatively high incidence –34, partA combination of anti-CTLA-4 and anti-PD1/PDL1 was found to be an independent risk factor of ICIs-associated AKI in several studies , 9, wherMultisystem irAEs could be induced during the treatment with ICIs, mainly attributed to the unrestricted activation of the immune system and the off-target mode of ICIs . The irAThere still be some limitations in our study. Firstly, all of the eligible studies were retrospective; thus, other possible confounding factors may influence the results. Secondly, a specific risk factor only reported in only one study was excluded in the present study due to it is impossible to pool the results in a meta-analysis. Thirdly, the relation between the factors with positive results and ICIs-related AKI could only be association. Further studies still need to be introduced to clarify the causal relationship. What’s more, reporting to FAERS was voluntary, hence the relationship between concurrent drugs  exposure in patients with ICIs and suspected adverse event (AKI event) was not clear and definite. Also, differences in the definition of AKI were unclear. Besides, detailed clinical information such as PD-1 status, patient’s baseline renal function, or other risk factors related to AKI were missing. Finally, comparisons of incidence for adverse events through a disproportionality analysis might be influenced by many confounding factors, such as reporting bias and lack of denominator data .Following the increasing use of ICIs, AKI related to these new anti-tumor agents have been reported and prompted more investigations. Based on meta-analysis of observational studies and real-world pharmacovigilance study of FAERS, our results suggested that drugs exposure , coexisting diabetes mellitus, genitourinary cancers, combination therapy of ICIs and extrarenal irAEs may increase the risk of AKI events in ICI-treated patients. Future studies are needed to investigate the mechanism and optimal management of ICIs-associated AKI.Additional file 1. Supplementary figure.Additional file 2. Supplementary tables.


Early fetal sex determination is worthy of providing alertness about possible x-linked disorders, as well as predicting sex-related pregnancy complications and outcomes. Satisfying the curiosity of parents is another advantage. In this way, several studies have been performed which have shown conflicting results.We planned a systematic review for identifying any plausible role of Fetal Heart Rate (FHR) for early predicting fetal sex during the first trimester of non-complicated pregnancies.This is a meta-analysis in which PubMed and Scopus databases were searched using different related keywords to find similar articles up to December 2022. Then the articles were screened to find eligible articles and finally, the articles entered in the meta-analysis were analyzed using Stata software . Standardized mean difference (SMD) and their 95% confidence interval (CI) were estimated.p = 0.012, I-squared = 69.0%). The results of meta-analysis with a random model showed that there is no significant difference between male and female genders in terms of mean FHR .A total of 223 articles were evaluated and five articles were included in the meta-analysis. The results showed that there is a significant heterogeneity between the articles (This systematic review and meta-analysis showed that even though male fetuses show faster FHR but such sex-related difference is minimal. Therefore, first-trimester FHR is not a reliable predictive test for fetal sex determination. Further studies are recommended to achieve a more precise conclusion.PROSPERO: CRD42023418291. Fetal sex determination is one of the integral parts of second-trimester sonograms, which is achievable by direct visualization of the fetus’s external genitalia  but earlAlthough laboratory exams including invasive methods like amniocentesis, chorionic villous sampling , 6, and Sex prediction based on Fetal Heart Rate (FHR) which can be achieved even by old sonographic devices, is another suggested simple method. Several studies have investigated the relationship between perinatal fetal heart rate and fetal gender. Results are inconsistent as the study by Hall et al. in 1993 showed female fetuses have FHR greater than 140 bpm  but mostA couple of studies also evaluated this relationship in the first trimester which continued to show different but non-significant results –19. FHR Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA)” , "Fetal Heart Rate"[tw], "Fetal sex"[tw], "Fetus sex"[tw], "Fetal sex"[all] and "Fetal Gender"[all]. All searches were conducted by one author (AAH). In addition to the mentioned databases, to find gray literature, Google Scholar was also searched and the title of found studies were screened up to page 20 (200 studies) of Google Scholar.In this meta-analysis, studies with the following characteristics were included in the analysis. 1) Articles whose study population is first-trimester pregnancies that underwent ultrasound and the results of the fetal heart rate in the first trimester were compared with the gender of the fetus that was determined in the second trimester. 2) Studies with full text in English, 3) and in terms of study design, all cross-sectional, case–control, and cohort articles can be included in this review. The study does not include case reports, case series, commentary, and letter to the editor. There is no time limit in search and screening. Finally, all the studies that reported the necessary data, including the mean and standard deviation of the fetal heart rate, along with the sample size by gender, were included in the analysis.The articles retrieved from the databases were entered into Endnote X8 software. First, duplicate studies were removed, and then the studies were screened based on the study title and abstract, and the studies that met the inclusion criteria were retained for further evaluations, and the rest were removed. After that, studies were screened using the full text of the remaining articles, and studies that did not meet the inclusion criteria were excluded. Finally, the required data were extracted from the remaining articles. Screening of articles was done by two researchers independently (SN and MHK), and in cases where two authors did not agree, a decision was made regarding the removal or keeping of the articles with the consultation of another author (FS).From the remaining articles that were candidates for meta-analysis, the required data were extracted by two authors independently (SN and MHK), and in cases where there was a disagreement between the authors, decisions were made in consultation with the other author (AAH). The data extracted from the studies included: the name of the first author, the year of publication of the study, the total sample size as well as the sample size by gender of the fetus, the mean age of mothers, the mean and standard deviation of the FHR by gender, country of study and study quality score.To assess the risk of bias or study quality, an assessment of the remaining studies was performed using the Newcastle–Ottawa Assessment Scale (NOS). This scale is designed for case–control, cohort, and cross-sectional studies. Using this checklist, articles are classified into three groups based on study quality, including articles with a score of 6 to 9 (low risk), articles with a score of 3 to 6 (moderate risk), and articles with a score of less than 3 (high risk). The items in this checklist include items related to the selection of participants, comparison, and exposure.metan” package was used in Stata software version 13 for data analysis . To check the heterogeneity among studies, in addition to the I-square statistic, the chi-square test was used, and in cases where there was significant heterogeneity between studies, the random-effects model was used to combine the results. Since the aim of the study was to compare the mean of a quantitative variable (FHR) between two groups , the effect size reported in the plot and the text is the standardized mean difference (SMD) ). Since the number of studies was relatively small, both the graphical method (funnel plot) and statistical method of Egger tests were used to investigate the possibility of publication bias, and based on the results of these methods, a decision was made regarding publication bias, and due to the publication bias, the trim and filled method was used. Finally, sensitivity analysis was also performed to check the impact of each study on the overall estimated SMD.Table and forest plot were used to report the results. “The steps of searching and screening the retrieved studies from the databases are shown in Fig. The oldest article was from 1989 in the USA  and the The standard Newcastle–Ottawa scale was used to evaluate the quality of each article. Based on this scale, all articles were classified as low risk.A total of 5 articles –19, 24 wAs it was presented in Fig. Investigation of the possibility of publication bias in the articles was done and the Funnel plot Fig.  showed aDue to the existence of publication bias, the trim and fill method was performed and the meta-analysis was repeated with a random-effects model  and the estimation made did not change significantly.For further investigations, a sensitivity analysis was also performed, in which each of the articles was excluded from the meta-analysis, and the meta-analysis was repeated with the remaining four articles. Different results were observed following the sensitivity analysis only when Oloyede et al. study  was remoThis systematic review and meta-analysis showed that even though male fetuses show faster FHR but such sex-related difference is minimal. Therefore, first-trimester FHR is not a reliable predictive test for fetal sex determination.Fetal sex determination is necessary in many ways. Some fetal disorders are x-linked and needed to be assessed during pregnancy . There iStudies of fetal cardiac function and fetoplacental circulation during the second and third trimesters of pregnancy revealed higher preload and lower afterload in male fetuses –31. SeveIn this meta-analysis, we evaluated 5 peer-reviewed published articles with a total population number of 4308 fetuses. Studies by Mckenna et al. , DuBose In assessing heterogeneity across studies, we faced a moderate degree of statistical heterogeneity that possibly could be related to different countries of origin (ethnic), mother’s age, gestational age, different sonographers experience, and accuracy of ultrasound machines. So, we utilized the random- Effects model instead of the fixed-effects model. In this line, the pooled estimate of the results exhibits no statistically significant difference in mean FHR between genders. The results of this meta-analysis showed that the mean FHR is similar between boys and girls. Therefore, it is not possible to predict the gender of the fetus by FHR.Thyroid hormone, glucocorticoids, and catecholamines are among the most important hormones which take part in the process of cardiac development, similarly in both sexes –39. WhatFetal heart variability is also different among genders so female fetuses exhibit greater heart rate dynamics in early gestational periods, suggesting that maturation of the cardiovascular system which is impressed by the autonomic nervous system, occurs earlier than that of males . One migUltrasonic determination of FHR can be obtained by M (Motion) mode and spectral Doppler analysis. Although the M mode is preferred over the latter, for a lower risk of thermal damage to the fetus , similarAcceptable population number is a strength of the current review, but there are also some limitations; Among our five eligible articles, the oldest one, considered pregnant women in any gestational age including the first trimester . Threate’s sonogram is not a dependable technique for early forecasting the fetal sex. The authors recommend further studies with a larger population for validation of the results.FHR determination in the first trimester


The potential effect of caffeine exposure during pregnancy on gestational hypertension (GH)/preeclampsia has attracted attention but remains unclear.A systematic literature search of PubMed, Embase, and Cochrane Library databases was performed until March 2022. Observational studies assessing the association between caffeine exposure during pregnancy and the risk of GH/preeclampsia were included. The study protocol was registered in PROSPERO: CRD42022322387.p = 0.800) and preeclampsia . Subgroup analyses comparing low to moderate doses with no/lowest doses showed that caffeine exposure during pregnancy was not significant associated with GH  or preeclampsia . Besides, subgroup analyses comparing high doses with no/lowest doses showed that caffeine exposure during pregnancy was not significant associated with GH  or preeclampsia .Ten studies involving 114 984 pregnant women (2548 diagnosed with GH and 2473 diagnosed with preeclampsia) were included. Comparing caffeine exposure with noncaffeine exposure, no significant association was found between caffeine exposure during pregnancy and the risk of GH (odds ratio [OR] = 0.99, 95% confidence interval [CI]: 0.90–1.08, This study found that caffeine exposure during pregnancy was not significantly associated with the risk of GH/preeclampsia. Hypertensive disorders of pregnancy (HDP) are one of the leading causes of maternal and perinatal morbidity and mortality worldwide, especially in low‐income and middle‐income countries.Caffeine is the most widely consumed psychoactive substance worldwide.It is estimated that approximately 70% of women in the United States consume caffeine continuously during pregnancy,We performed a systematic search on PubMed, Embase, and Cochrane Library databases from inception to 11 March 2022. The search items included  AND  AND . Moreover, the references of the included articles were manually checked for additional sources. The entire review process was mapped using the Preferred Reporting Items for Systematic Reviews and Meta‐analyses (PRISMA) flow diagram and the study protocol was registered in PROSPERO: CRD42022322387.1: were observational studies 2; assigned participants consumed caffeinated beverages to the exposure group, and participants consumed no or the lowest dose of caffeinated beverages to the nonexposure group3; reported risk estimates for the association between caffeine exposure during pregnancy and the risk of GH/preeclampsia4; were published in English.The meta‐analysis included studies with the following criteria, which1: pregnant women with a medical history of hypertension and/or renal disease, history of HDP in previous pregnancies, and history of diabetes mellitus and gestational diabetes mellitus were included.2 The articles were designed to investigate the association between caffeine exposure before pregnancy and the risk of GH/preeclampsia.3 No exactable data was available.4 When more than one paper was published by the same author or institution, the paper with the highest quality was selected.The exclusion criteria were as followsTwo researchers independently evaluated the eligibility of each study and exploited the desired information from it. Disagreements were arbitrated by a third researcher. The Newcastle–Ottawa Quality Assessment Scale (NOS) was used to assess the methodological quality of the included observational studies. Additionally, the following data were extracted: author, year of publication, country, study design, number of participants, time to recruit participants, pregnancy stage, outcome, caffeine intake sources, dietary assessment tool, and adjusted variables.p < 0.05 estimated as significant. OR and 95% CIs were applied as the effect sizes of caffeine exposure during pregnancy on the risk of GH/preeclampsia. As the dose and source of caffeine intake differed, a random effects model was used to combine effect sizes. Moreover, statistical heterogeneity among studies was assessed based on I2 statistics. The values I2 of 25%–50%, 50%–75%, and >75% were considered as low, moderate, and high heterogeneity, respectively.All statistical analyses were performed using Stata software version 12.0, with The flow chart of study selection is summarized in Figure The NOS consists of eight items and is divided into three dimensions, according to selectivity, comparability, and the study type‐outcome (cohort study) or exposure (case–control study).Since coffee/caffeine consumption was reported on various scales, we converted exposure data to a uniform measurement (mg/day). The amount of caffeine consumed per cup of coffee ranged from 74 to 107 mg, for tea 55 to 65 mg, for chocolate 16 to 56 mg and for soft drinks 24 to 95 mg.I2 = 51.0%, p = 0.800)    and soft drink  consumption during pregnancy was associated with an increased risk of preeclampsia, although the latter was not statistically significant . Moreover, pooled results of further subgroup analyses showed that consumption of chocolate  and soft drinks  during pregnancy was not significantly associated with the risk of preeclampsia  at high doses versus no/lowest doses. Additionally, pooled results of further subgroup analyses showed that consumption of coffee , chocolate , and soft drinks  during pregnancy was not significantly associated with the risk of preeclampsia  recommended that pregnant women consume up to two cups of moderate‐strength coffee per day.1 and A2A receptors.Although, as stated above, our results do not allow to draw a conclusion about causality, it is reasonable to speculate about possible mechanisms for the putative adverse effects of coffee or soft drink consumption on preeclampsia. Coffee contains several compounds such as kahweol, cafestol, chlorogenic acid, and trigonelline. These compounds have bidirectional effects on blood pressure regulation.The strengths of this article are that, to our knowledge, no meta‐analysis has examined the association between caffeine exposure during pregnancy and the risk of GH/preeclampsia. In addition, comprehensive subgroup analyses were established to explore the heterogeneity and compare the potential differences between different subgroups.However, there are also some limitations. First, the number of articles included in this meta‐analysis was limited. Second, there was no standardization of caffeine in estimating caffeine consumption, which could lead to measurement bias. Third, all eligible articles were observational studies that had certain biases, such as recall bias and selection bias. Fourth, some studies did not control for covariates or ignored some potential residual confusions. Taking race as an example, although a large prospective cohort study has noted significant racial differences in HDP,The results of this meta‐analysis showed that caffeine exposure during pregnancy is not significantly associated with the risk of GH/preeclampsia, regardless of the dose or source of caffeine intake. Moreover, more appropriately designed, larger‐scale studies are needed to explore the impact of caffeine exposure during pregnancy on the risk of GH/preeclampsia.Each author contributed significantly to concept and development of the present paper. Saisai Ni and Mengting Zhang designed the research process. Bangsheng Chen, Yujing He, and Yuexiu Si searched the database for corresponding articles and extracted useful information from the articles above. Yetan Shi and Ke Jiang used statistical software for analysis. Jingyi Shen and Jiaze Hong drafted the meta‐analysis. All authors had read and approved the manuscript and ensured that this was the case.This study was supported by the Natural Science Foundation of Ningbo, China (project No. 2018A610398), the Medical Health Science and Technology Project of Zhejiang Provincial Health Commission, China, (project No. 2022KY343), and the Science and Technology Program for Public wellbeing of Ningbo, China (project No. 2022AS069).The authors declare there was no conflict of interest.Table S1: Quality assessment of included studies in this meta‐analysis.Click here for additional data file.