Hesperidin is a naturally occurring bioactive compound that may have an impact on cardiovascular disease risks, but the evidence is not conclusive. To investigate further, this study aimed to explore the effects of hesperidin supplementation on cardiovascular risk factors in adults. A comprehensive search was conducted up to August 2022 using relevant keywords in databases such as Scopus, PubMed, Embase, Cochrane Library, and ISI Web of Science for all randomized controlled trials (RCTs). The results showed that hesperidin supplementation had a significant effect on reducing serum triglyceride (TG), total cholesterol (TC), low-density cholesterol (LDL), tumor necrosis factor-alpha (TNF-α), and systolic blood pressure (SBP), whereas weight was increased. However, no significant effect was observed on high-density cholesterol (HDL), waist circumference (WC), fasting blood glucose (FBG), insulin, homeostatic model assessment for insulin resistance (HOMA-IR), C-reactive protein (CRP), interleukin-6 (IL-6), body mass index (BMI), and diastolic blood pressure (DBP). The study also found that an effective dosage of hesperidin supplementation was around 1,000 mg/d, and a more effective duration of supplementation was more than eight weeks to decrease insulin levels. Furthermore, the duration of intervention of more than six weeks was effective in decreasing FBG levels. ApproxiPolyphenols, a large group of natural-based bioactive compounds, consist of three main classes, including flavonoids and non-flavonoids , 21. Manin vitro and in vivo studies. While there is great potential to leverage bioactive components of plants for the discovery and development of modern therapies, the results from randomized controlled trials (RCTs) examining the effect of hesperidin supplementation on the constellation of risk factors for CVD risk factors  are inconclusive due to high variability between human trials. Also, the addition of new articles about CVD risk factors, as well as a complete assessment of all CVD risk factors, is the reason for this systematic review. Therefore, this study aimed to evaluate the effect of hesperidin supplementation on cardiovascular disease risk factors in adults.Research has shown that hesperidin treatment can affect CVD risk factors, such as insulin resistance , hypergl2.PRISMA declaration was used in this study for reporting preferred reporting items for systematic reviews and meta-analyses .2.1.PubMed/Medline, Scopus, Web of Science, EMBASE, and the Cochrane databases, as well as Google Scholar, were searched to identify available RCT on hesperidin supplementation and CVD risk factors published up to August 2022. Additionally, research bibliographies were reviewed to identify any potential missing studies. Neither the time nor the language of the publications were restricted. We used the PICO  search framework to search for items related to hesperidin supplementation and CVD risk factors and to search for items related to hesperidin supplementation and all the outcomes. We used relevant formatting for each folder. We used a combination of MeSH terms and keywords as the first step of the data collection process, to ensure that no studies were missed. The reference lists for all related studies were manually searched using the following keywords: (“hesperidin” OR “hesperitin” OR “citrus flavonoid” OR “orange juice” OR “citrus flavanones” OR “orange polyphenols”) AND .2.2.in vitro. Additionally, we excluded gray literature, conference abstracts, editorial papers, books, and RCTs without a placebo or control group. Similarly, studies with the combination of hesperidin with vitamins or minerals, without a concurrent placebo-controlled group, or that reported mean CVDs to risk factors  as the only outcome were excluded.Research studies were included if they met the following criteria: (1) RCTs ; (2) oral intake of hesperidin; (3) assessed the effects of hesperidin supplementation on CVD risk factors; (4) RCTs with two or more eligible arms were considered as separate studies; (5) study participants were adults (≥18 years old); (6) provided means and standard deviations (SDs) for hesperidin, provided means and SDs for CVD risk factors or any other effect sizes from which the calculation of mean and SD was possible. No language restrictions were imposed on the searches, and only human studies were included. Two authors (ASK and OA) independently assessed the validity of the qualifying studies by assessing their titles and abstracts, extracting results, and investigating the validity of the included publications. The following studies were excluded: animal studies, reviews, and studies conducted 2.3.The authors ASK and OA independently re-checked all eligible RCTs and extracted the following information. Several data were extracted for further analysis, including the first author’s name, country, year of publication, type of clinical study, characteristics of participants , randomization, blinding, sample size, dose, and forms of supplemented hesperidin, duration, and related information. We collected the mean and SD for CVD risk factors at the start and end of each intervention . Whenever no such data were available, the mean difference was calculated by subtracting the mean value of the baseline from that of the endpoint. If the hesperidin doses were reported in g/day, we converted them to mg/day.2.4.Authors ASK and OA separately assessed the quality of the work, with any differences settled by discussion. We used the Cochrane Collaboration tool to assess the quality of the studies . A serie2.5.p-values were less than 0.05; all tests performed were two-tailed. We calculated the pooled weighted mean difference (WMD) using a random effects model developed by Der Simonian & Laird, which was based on the existing heterogeneity. This was due to the different intervention doses, participant health, sample sizes, ethnicity, and length of intervention (2 + (SD at the end of study)2 − (2r × SD at baseline ×SD at the end of study)] (2) statistic (p < 0.05 and I2 > 40%) after visual inspection of forest plots or Cochrane’s Q test  and obese (≥30 kg/m2)], age .We used Stata version 11.0  to conduct this meta-analysis. We considered all tests to be statistically significant when the rvention . We calc study)] . Mean dich group . We usedch group . Heterogs Q test . A signiogeneity . SubgrouThe eager test and the funnel plot test were implemented to assess publication bias in studies evaluating the effects of hesperidin supplementation on CVD risk factors . We cond2.6.GRADE  was used to assess and summarize the overall certainty of evidence across the studies .3.3.1.n = 338) and irrelevant research indicated by titles and abstracts , 27 possibly relevant articles were taken into account for a full-text review. After the screening, 14 studies were excluded due to not reporting the desired data. All of the studies were in the English language. Finally, 13 papers in total were included  using the Cochrane risk of bias assessment method , 48–57. 3.5.p = 0.042) . Hesperidin supplementation significantly decreased TG at doses >500 mg/d  and duration >6 weeks , according to subgroup analysis (p = 0.004), age < 50 years  and baseline TG >150 , as well as in studies conducted in both sexes , generally showed a substantial decline after hesperidin supplementation  , accordi= 0.042) , 54–57. analysis . The subentation .3.6.p = 0.023)  (p = 0.021), BMI > 30 , and age < 50 years  according to the subgroup analysis. Moreover, hesperidin significantly reduced TG at doses of >500 mg/d  and duration of more than 6 weeks  (p = 0.037)  , with no= 0.769) . Hesperi= 0.039) , and in = 0.037) .3.7.p = 0.017) . The reduction in LDL after hesperidin supplementation was significant at doses greater than 500, according to the subgroup analysis based on the intervention dose . Other subgroup analyses also showed that hesperidin significantly reduced the LDL in non-CVD patients , individuals with BMI > 30 , age < 50 years , and in studies conducted in both sexes   , with no= 0.027) .3.8.p = 0.157) . Our subgroup analyses showed no significant between-study heterogeneity in all subgroups except in studies conducted in patients with CVD   , and the = 0.04) .3.9.p = 0.110) . Furthermore, subgroup analysis showed that hesperidin supplementation affected FBG  when a high dose of hesperidin (>500 mg/d) was used and the duration of supplementation was longer than 6 weeks  in individuals with CVD , age < 50 years , and in a patient with baseline FBG ≥100 mg/dL  .In total, 430 participants from nine studies were included in the analysis , 54–57. = 0.110) . Between= 0.001) . Subgrou3.10.p = 0.145)   , and the= 0.830) .3.11.p = 0.180) . Subgroup analyses showed that hesperidin supplementation significantly affected HOMA-IR in participants aged ≥50 years  and in the low dose interventions (≤500 mg/day)   . Between= 0.039) .3.12.p = 0.560)  , those that used >500 mg hesperidin , studies that enrolled obese patients , studies conducted in both sexes , and when patients were aged <50 years .The overall analysis of nine studies enrolling 433 participants , 57 indi= 0.560) , with hi= 0.002) . Between3.13.p = 0.121) . In subgroup analysis, we observed that hesperidin significantly reduced IL-6 at doses of >500 mg/d  and in patients with CVD   . The bet= 0.013) .3.14.p = 0.012) . After subgroup analysis, we found that between-study heterogeneity disappeared in studies with intervention dose >500 , studies conducted in both sexes , and studies that enrolled obese  and overweight  individuals. Additionally, subgroup analyses revealed that hesperidin supplementation significantly reduced TNF-α in every subgroup except for those who received an intervention dose of ≤500 , those who were of normal weight (18.5 < BMI < 24.9) , and in studies conducted in men only   . Between; p = 1) .3.15.p < 0.001) by combining the data from these investigations . Subgroup analyses revealed that hesperidin supplementation had a significant effect on weight in participants aged ≥50 , in the high dose interventions (>500) , in trial duration ≤6 weeks , and in patients with CVD   .3.16.2, 95%CI: −7.45, 2.66; p = 0.35)   . Between< 0.001) .3.17.p = 0.051). The test for between-study heterogeneity was not significant   .3.18.p = 0.046) (p = 0.055), overweight patients (25 < BMI < 29.9) , and those with a duration of intervention ≤6 weeks   , with no= 0.009) .3.19.p = 0.415)    and the = 0.172) .3.20.non-linearity = 0.004; non-linearity = 0.001; non-linearity = 0.001; non-linearity = 0.551; non-linearity = 0.173; non-linearity = 0.221; non-linearity = 0.341; non-linearity = 0.313; non-linearity = 0.447; non-linearity = 0.233; non-linearity = 0.0.231; non-linearity = 0.696; non-linearity = 0.502; non-linearity = 0.616; non-linearity = 0.501; non-linearity = 0.248; We utilized a non-linear dose–response regression to analyze the dose–response relationship between hesperidin supplementation and cardiovascular risk variables. Hesperidin dose and insulin levels were shown to have a non-linear relationship  (p = 0.032) (p = 0.009)    and dura= 0.032)  and chan= 0.009) . There w3.22.Our sensitivity analysis revealed that the effect sizes for the impact of hesperidin on TG, TC, LDL, HDL, insulin, HOMA-IR, CRP, IL-6, BMI, weight, SBP, and DBP remained robust even after removing each study one by one. While based on the results of sensitivity analysis, data on FBG was sensitive to studies by Salden et al.  WMD = − (WMD = −3.23.3.24.p = 0.052) (p = 0.159) (p = 0.365) (p = 0.463) (p = 0.492) (p = 0.307) (p = 0.764) (p = 0.343) (p = 0.968) (p = 0.274) (p = 0.486) (p = 0.007) (p = 0.035) (p = 0.005) (p = 0.001)  , LDL (p = 0.159) , FBG (p = 0.365) , insulin= 0.463) , HOMA-IR= 0.492) , CRP (p = 0.307) , IL-6 (p= 0.764) , BMI (p = 0.343) , WC (p == 0.968) , SBP (p = 0.274) , and DBP= 0.486) ; however= 0.007) , HDL (p = 0.035) , weight = 0.005) , and TNF= 0.001) .4.We conducted a comprehensive systematic review and meta-analysis to examine the effects of hesperidin supplementation on cardiovascular risk factors in adults. Our analysis investigated the impact of hesperidin on various biomarkers associated with CVD risk, including the lipid profile, inflammatory markers, blood glucose and insulin, blood pressure, and weight. While some previous systematic reviews have examined the effects of hesperidin, none have analyzed all of these biomarkers, and the results have not been conclusive. Our analysis revealed that hesperidin supplementation significantly reduces TG, TC, LDL, TNF-α, and SBP, while also causing weight gain in participants.have positive effects on cardiovascular health , heart fatty acid-binding protein (H-FAB), and cutaneous fatty acid-binding protein (C-FAB) expression. This ultimately results in an improvement in hypercholesteremia and fatty liver in animal models, such as rats  has on chronic disease biomarkers in humans over 18 years of age and found results that were not conclusive about HDL or LDL . In 2019 as rats . Hesperi as rats , 66. HMG as rats . A study as rats . Hesperi as rats . Hesperi as rats . Further as rats . These mOur meta-analysis also revealed that the efficacy of hesperidin in reducing TG levels was influenced by certain factors. Subgroup analysis indicated that a longer intervention duration of more than six weeks and a daily dose of more than 500 mg were associated with a greater reduction in TG levels. Additionally, the supplementation of hesperidin was found to be more effective in reducing TG levels in obese subjects and participants younger than 50 years old. Our findings also suggest that hesperidin is more effective in reducing TG levels when the baseline TG of participants is higher than 150 mg/dL. The bioavailability of hesperidin in the blood of participants may be affected by their microbiome, as colon flora can convert hesperidin into insoluble metabolites that cannot be efficiently absorbed . AlthougThis meta-analysis showed supplementation of hesperidin had no effects on FBG and HOMA-IR when pooled effects sizes were included. A systematic review conducted by Tadros et al. in 2021 to evaluate the impact that hesperidin in 100% orange juice had on chronic disease biomarkers showed inconclusive results of its effect on insulin . In 2020Our analysis also revealed a non-significant dose–response effect of hesperidin dosage on the reduction of serum FBG levels, with a gradually decreasing trend observed from 300 to 1,000 mg of hesperidin per day. This suggests that hesperidin may exert its FBG-lowering effect through a wide range of dosages. We also found a significant U-shaped duration response of hesperidin supplement duration on FBG levels. While a 3- to 4-week intervention increased FBG levels, longer interventions of more than 6 weeks up to 12 weeks significantly reduced FBG levels. These findings suggest that hesperidin supplementation for more than 6 weeks may be the most effective in reducing FBG levels.In addition, our non-linear dose–response analysis revealed a significant association between hesperidin supplementation dose and insulin levels. There was a significant dose–response effect of hesperidin on insulin levels, with insulin levels gradually decreasing with hesperidin in the range of 300 mg/day to 1,000 mg/day. Moreover, 1,000 mg/day of hesperidin was found to be the most effective in lowering insulin levels. The duration-response analysis also found a significant U-shaped duration-response effect of hesperidin on insulin levels. Insulin levels showed an increasing trend when the duration of intervention was between 3 to 8 weeks, while a decreasing trend was seen when the duration of intervention was more than 8 weeks. These findings suggest that longer interventions of more than 8 weeks may be needed to observe significant reductions in insulin levels with hesperidin supplementation.The antidiabetic effects of hesperidin have been demonstrated in animal models, with hesperidin exerting its effects by stimulating insulin secretion, stimulating glucose uptake in peripheral tissue, inhibiting gluconeogenesis (downregulation of glucose-6-phosphatase), and activating glycolysis (upregulation of glucokinase) . HesperiThis meta-analysis indicated that hesperidin supplementation significantly lowered the TNF-α level in serum. A systematic review conducted in 2021 that evaluated the impact hesperidin in 100% orange juice had on chronic disease biomarkers reported that hesperidin in orange juice decreased IL-6 and TNF-α compared to the control group . ReductiThis meta-analysis indicated hesperidin supplementation significantly lowered the TNF-α level in serum. However, it did not affect the TNF-α level when a low dose (<500) was used and in studies conducted on men only or normal-weight participants. A low dose of hesperidin supplementation may not be sufficient to reduce TNF-α levels due to its low bioavailability . It shouHesperidin, as a flavonoid, seems to have anti-inflammatory properties. Inflammation is a complex process that includes many inflammatory mediators such as iNOS (inducible nitric oxide synthase) and cyclooxygenase-2 (COX-2), which enhance the nitric oxide level and prostaglandins. Several cytokines are also related to the inflammatory process and are elevated in these processes, including TNF-α, IL-1β, and IL-6. These mediators and cytokines are regulated by transcription factors, of which kappa-light-chain-enhancer of activated B cells (NF-κB) has the central role. NF-κB signaling pathway also activates mitogen-activated protein kinases (MAPKs), which are involved in inflammatory processes . Many stThis meta-analysis indicated that hesperidin supplementation did not affect BMI and WC but an increasing effect was detected on weight. In a systematic review conducted by Pla-Paga et al. in 2019, hesperidin did not affect weight or BMI . Our resOur meta-analysis revealed that hesperidin induced weight gain in participants, specifically in those with intervention durations of less than 6 weeks, in those with intervention doses of more than 500 mg per day, in participants older than 50 years old, and in participants with a CVD history. Increasing intervention dose and duration improved hesperidin effects probably by increasing its bioavailability in the blood . HoweverOur findings indicated that hesperidin supplementation decreased SBP in overall effect; however, did not affect DBP. Pla-Paga et al. conducted a systematic review in 2019 to evaluate the effects of hesperidin consumption on cardiovascular risk biomarkers in animal studies and human randomized clinical trials. Body weight, BMI, body fat, SBP, DBP, glucose level, insulin, TC, LDL, HDL, and TG were evaluated in human studies, but no significant changes were found in any of them . In addiSubgroup analysis showed that hesperidin supplementation is more effective in reducing SBP when the duration of the intervention is shorter than six weeks or when participants are overweight (BMI:25–29.9). Although it is not clear why a shorter intervention duration is more effective in reducing SBP, it could be due to the compensating mechanisms that activate in the long term and reduce the effect of hesperidin. Furthermore, different hesperidin metabolites, which have various extents of hypertensive effects, should be considered when interpreting research results .in vitro studies that showed that hesperidin inhibits mast cell inflammatory responses and inflammatory cytokine secretion and decreases TNF-α activity (Mechanistic evidence of the effect of hesperidin on CVD risk factors has been investigated extensively in animal models but to less extent in human subjects; however, its mechanism of action is not entirely discovered. Hesperidin possesses anti-inflammatory and anti-oxidative activities and also affects lipid and glucose metabolism. Understanding the effects of hesperidin and its mechanism of action is of great importance for using this substance as a treatment for various diseases. The anti-hyperglycemic effect of hesperidin has been investigated in mice. Hesperidin exerts its effects on FBG by modulating the activity of key enzymes in glucose metabolism. Hesperidin maintains glucose metabolism by regulating PPAR-c, a nuclear transcription factor . Hesperiactivity , 93. Hesactivity . Hesperiactivity . Dose-deactivity . AlthougThis meta-analysis has some limitations and strengths that should be addressed. Participants in different studies had varying degrees of health statuses, different mean ages, and different BMIs, which affected the heterogeneity of the data. Different study designs could lead to heterogeneity in results and should be considered seriously. Heterogeneity in the included population, hesperidin dosage, and intervention duration was seen in studies. Hesperidin can be found in consumed food and beverages, but its bioavailability is affected by the food matrix in which it is consumed, so controlling the diet of participants is of great importance. However, not all studies vigorously monitored dietary intake . Further5.This meta-analysis indicated that hesperidin supplementation had a lowering effect on TG, TC, and LDL serum levels, and it also lowered TNF-α and blood pressure. However, further well-designed RCTs and mechanistic studies are needed to elucidate the effect of hesperidin on CVD risk factors, specially FBG, insulin resistance, blood pressure, HDL, and inflammatory markers. It seemed that the effective dosage and duration of hesperidin supplementation for decrement of insulin level are approximately 1,000 mg/d and more than 8 weeks. In addition, the findings revealed a non-linear association between the duration of hesperidin intervention and FBG, with a decrease in FBG levels appearing after 6 weeks of hesperidin consumption.The original contributions presented in the study are included in the article/AK designed the study. AK and OA developed the search strategy, extracted the data, conducted the analyses, and assessed the risk of bias of the meta-analyses. NR, SF, and FG drafted the manuscript. FS, OA, and AK interpreted the results. FS, SK, and OA revised the manuscript. All authors read and approved the final manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


Purpose: Tacrolimus (Tac) is a widely used immunosuppressive agent in kidney transplantation. Cytochrome P450 (CYP), especially CYP3A4 enzymes are responsible for the metabolism of drugs. However, the correlation between plasma Tac concentration and CYP3A4*22 gene variants is controversial. This meta-analysis aims to evaluate the association between CYP3A4*22 polymorphism and the dose-adjusted trough concentration (C0/D) of Tac in adult kidney transplant patients.Methods: We conducted a literature review for qualifying studies using the PubMed, Web of Science, and Embase databases until July 2023. For the continuous variables (C0/D and daily dose), mean difference (MD) and corresponding 95% confidence intervals (CIs) were calculated to evaluate the association between the CYP3A4*22 and Tac pharmacokinetics. We performed an additional analysis on the relationship of CYP3A5*3 with Tac PKs and analyzed the effects of CYP3A4*22 in CYP3A5 non-expressers.Results: Overall, eight eligible studies with 2,683 renal transplant recipients were included in this meta-analysis. The CYP3A4*22 allele was significantly associated with a higher C0/D  and lower mean daily dose requirement . An additional meta-analysis demonstrated that carrying the CYP3A5*3 polymorphism greatly impacted Tac blood concentration. From the result with CYP3A5 non-expressers, CYP3A4*22 showed significant effects on the Tac C0/D and dose requirement even after adjusting the effect of CYP3A5*3.Conclusion: Patients with CYP3A4*22 allele showed significantly higher plasma C0/D of Tac and required lower daily dose to achieve the therapeutic trough level after kidney transplantation. These findings of our meta-analysis may provide further evidence for the effects of genetic polymorphism in CYP3A4 on the PKs of Tac, which will improve individualized treatment in a clinical setting. Tacrolimus (Tac) is a widely used maintenance immunosuppressive agent to prevent graft rejection in kidney transplantation. Tac suppresses T-cell activation by inhibiting the calcineurin activity and exhibits excellent graft survival with a low incidence of rejection . HoweverTac is also characterized by its high inter-individual variability in its pharmacokinetics (PKs) . This maCYP3A5*3  is the most significant genetic determinant of Tac PKs  3A4 and 3A5, genetic polymorphisms related to the expression of these proteins have been studied to explain the between-subject PK variability . Among t Tac PKs . This poe allele . Accordie allele .CYP3A4 polymorphisms could influence the PKs and efficacy of related drugs , a novel variant of CYP3A4, has been reported to have low messenger RNA (mRNA) expression and low activity of CYP3A4; accordingly, its relationship with drug response has been widely studied  AND (Tacrolimus OR FK506) AND (CYP3A4*)  guidelines . A comprCYP3A4*) .CYP3A4*22 genotypes and trough concentrations (C0) of Tac at steady state; 4) adjusted C0 by daily dose; and 5) expressed data as the mean with standard deviation (SD) or the median with range.The eligible studies were selected in the analysis if they 1) were cohort studies; 2) involved the adult renal transplantation patients who took tacrolimus; 3) evaluated the association between in vitro or in vivo studies; or 3) were conducted on patients with a single dose of Tac. If there was possibility of data overlap among the studies, only the most recent and comprehensive data was included.The studies were excluded if they were 1) not original articles ; 2) 0 of Tac, the allele frequency of CYP3A4*22, dose-adjusted trough concentration (C0/D), and the daily dose of Tac according to post-transplantation period.Two authors (JSK and SS) performed the initial screening independently using Endnote to exclude duplicate studies. Next, the list of studies was compared, and consensus was achieved through discussion. Subsequently, both reviewers (JSK and SS) independently assessed the titles and abstracts, excluding studies that did not meet the inclusion and exclusion criteria. Throughout this process, the reviewers ensured methodological consistency and error reduction in the extraction techniques. Then, both authors (JSK and SS) independently evaluated the full text of all relevant studies to determine their eligibility. All studies that did not meet the eligibility criteria during the second screening were documented, along with the reasons for their exclusion. In case of any disagreements on study selection, a consensus was reached through discussion with a third reviewer (JY). For each study, extracted data were as follows: first authors, publication years, study design, country, ethnic background, characteristics of participants , immunosuppressive protocol, alleles studied, genotyping methods, measurement methods for the C0/D was calculated by the plasma trough concentration (ng/mL) of Tac divided by daily dose (mg), expressed as ng/mL per mg  system was adopted to rate the quality of the evidence. The total score of NOS ranges from 0 to 9; 0–4 points were assigned for the selection of the population, 0–2 points for comparability, and 0–3 points for the outcomes. For comparability, 1 point each was awarded if studies matched or adjusted with the age or other known risk factors.0/D and daily dose), mean difference (MD) and corresponding 95% confidence intervals (CIs) were calculated to evaluate the association between the CYP3A4*22 and Tac PKs.For the continuous variables  was used; if not, the random-effect model (DerSimonian-Laird method) was applied  version 5.4  and R software (version 3.6.0). A  applied . Begg’s  applied .CYP3A5*3 with Tac C0 and daily dose. In order to observe the independent influence of CYP3A4*22 while controlling for CYP3A5*3, we analyzed the effects of CYP3A4*22 in CYP3A5 non-expressers.Sensitivity analyses were conducted to evaluate the robustness of the results by omitting the factor to assess its influence on the overall estimate. The first sensitivity analysis was performed by excluding each post-transplantation period at a time sequentially, and another sensitivity analysis by omitting studies that scored lower than 7 on the NOS system. We performed an additional analysis on the relationship of n = 7), did not adjust concentration by dose (n = 6), did not investigate CYP3A4*22 (n = 4), could not express data in mean with SD (n = 1), and administered a single dose (n = 1). Finally, eight cohort studies were selected, including data of 2,624 patients in the meta-analysis. The main characteristics of the eligible studies are presented in A total of 950 studies were retrieved by the literature search and 475 duplicates were removed . After eCYP3A4*22 genetic polymorphism on C0/D were evaluated by meta-analysis . Except for the first 2 weeks post-transplantation, statistically notable differences in the C0/D of Tac were detected according to CYP3A4*22 genotypes. Although substantial heterogeneity across the studies was found , no subgroup difference was reported among the six different time periods (p = 0.78). Begg’s and Egger’s tests indicated no evidence of publication bias .The effects of the analysis  . Except for 1-year post-transplantation, significant differences in the daily dose were observed between CYP3A4*22 carriers and CYP3A4*1/*1 carriers. Similar to C0/D, there were substantial heterogeneity  but no subgroup significant difference (p = 0.49). Results from Begg’s and Egger’s tests indicated no statistical evidence of publication bias .Six studies  were anae of Tac . When da0/D, heterogeneity was mitigated when data measured in the first week after transplantation . The results of the dose requirement showed an MD range of −2.19 to −1.85 mg/day with an I2 range of 15%–80%. For dose requirements, heterogeneity was greatly reduced when data measured 1 year after transplantation . Another sensitivity analysis was performed with the studies that scored 7 or higher on the NOS system . This finding indicates that the CYP3A4*22 and CYP3A5*3 polymorphisms are independently associated with Tac exposure. Tac PKs . An addi studies  and GBR/0/D, the heterogeneity was reduced when data collected within the first week after transplantation was excluded. This suggests that the observed heterogeneity may be attributed to the early post-transplant period, which is characterized by the insufficient function of the transplanted graft.This meta-analysis revealed a significant degree of heterogeneity among the included studies. Statistical heterogeneity can be attributed to the small number of included studies and the wide range of sample sizes, varying from 52 to 1,366 patients. Also, some factors that can potentially contribute to clinical heterogeneity, including variances in analytic methods and target trough concentration. In the sensitivity analysis of CCYP3A4*22 is rarely found in African or Asian descent, whereas the allele frequency of CYP34*22 is approximately 8% in Caucasians (This study has several limitations. First, all included studies were conducted in European populations. This was because ucasians . Second,CYP3A4*22 allele carriers showed significantly higher plasma C0/D of Tac and required a lower daily dose to achieve the therapeutic trough level after kidney transplantation. These findings of our meta-analysis may provide further evidence for the effects of genetic polymorphism in CYP3A4 on the PKs of Tac, which will improve individualized treatment in a clinical setting.


There was also found a significant correlation between PTx and NEC . As we were not able to reduce heterogeneity in the assessment of the relationship between PTx and IVH, no conclusion could be taken.Platelet transfusions (PTx) are the principal approach for treating neonatal thrombocytopenia, a common hematological abnormality affecting neonates, particularly preterm infants. However, evidence about the outcomes associated with PTx and whether they provide clinical benefit or harm is lacking. The aim of this systematic review and meta-analysis is to assess the association between PTx in preterm infants and mortality, major bleeding, sepsis, and necrotizing enterocolitis (NEC) in comparison to not transfusing or using different platelet count thresholds for transfusion. A broad electronic search in three databases was performed in December 2022. We included randomized controlled trials, and cohort and case control studies of preterm infants with thrombocytopenia that (i) compared treatment with platelet transfusion vs. no platelet transfusion, (ii) assessed the platelet count threshold for PTx, or (iii) compared single to multiple PTx. We conducted a meta-analysis to assess the association between PTx and mortality, intraventricular hemorrhage (IVH), sepsis, and NEC and, in the presence of substantial heterogeneity, leave-one-out sensitivity analysis was performed. We screened 625 abstracts and 50 full texts and identified 18 reports of 13 eligible studies. The qualitative analysis of the included studies revealed controversial results as several studies showed an association between PTx in preterm infants and a higher risk of mortality, major bleeding, sepsis, and NEC, while others did not present a significant relationship. The meta-analysis results suggest a significant association between PTx and mortality (RR 2.4, 95% CI 1.8–3.4;     Conclusion: Platelet transfusions in preterm infants are associated to a higher risk of death, sepsis, and NEC and, possibly, to a higher incidence of IVH. Further studies are needed to confirm these associations, namely between PTx and IVH, and to define the threshold from which PTx should be given with less harm effect.The online version contains supplementary material available at 10.1007/s00431-023-05031-y. This condition mainly affects preterm infants and critically ill neonates and it occurs due to decreased platelet production or increased consumption, or both  than in the non-transfused group. Significant heterogeneity was present across the studies , so a leave-one-out sensitivity analysis was conducted to identify potential sources of heterogeneity in the quantitative analysis. Regardless of which study is removed, there is always a significantly higher risk of death in the transfused group (Supplp < 0.0001).Among the studies that compared preterm infants who received PTx with those who did not, five , 21, 22 oup Suppl. HeterogRegarding the evaluation of different transfusion thresholds, PlaNeT-2 Trial –27 foundp < 0.001), with odds ratio (OR) of 4.3 for developing any grade of IVH (95% CI: 2.3–8.2) and 41 for grade III or IV (95% CI 14.0–146.6). Andrew et al. . Due to significant heterogeneity across the studies , a leave-one-out sensitivity analysis was conducted (Supplp < 0.0001).Despite that, we conducted a meta-analysis for both outcomes using data present in several studies , 21, 22.p < 0.0001), with no heterogeneity detected .The forest plot for NEC, illustrated in Fig. The qualitative analysis of the studies included in this systematic review has revealed the controversial nature of the topic and the lack of consistent evidence about the effects of PTx in preterm infants. In fact, several studies have shown an association between PTx in thrombocytopenic preterm infants and a higher risk of mortality, major bleeding, sepsis, and NEC, while others did not show a significant relationship. The same happens with studies assessing different thresholds . These discrepancies may be partly attributed to methodological differences across the studies . These methodological disparities pose challenges when comparing the studies and drawing definitive conclusions.Our meta-analysis aimed to assess the potential relationships between PTx and the outcomes in study. Despite the presence of high heterogeneity, significant risk-associations were found between PTx and adverse outcomes. A leave-one-out sensitivity analyses for mortality and sepsis outcomes allowed for decreased heterogeneity. Motives of such heterogeneity are already mentioned above. Based on these results, PTx in preterm infants are associated to increased risk of mortality, sepsis, and NEC. Evidence regarding the association between PTx and IVH is too heterogeneous and a meta-analysis could not be conducted.9/L) resulted in lower mortality rates and hemorrhage episodes in infants than transfusions at a higher threshold (PC below 50 × 109/L). The liberal transfusion group received more PTx and had more adverse events than the restrictive one. Another RCT was conducted by Kumar et al. , may be a consequence of PTx rather than the primary cause. Likewise, it is possible that a higher number of PTx are administered due to the presence of IVH or other hemorrhage, rather than bleeding being caused by the transfusions. To minimize these confounding effects, in every included study, we ensured that the outcomes were not present before the intervention. However, the non-randomized design of some studies introduces the potential for selection bias, whereby infants who received PTx may have had higher levels of morbidity, which could have influenced the occurrence of adverse outcomes. Additionally, there is missing data from the original studies regarding the index of severity of illness and causes of death. Furthermore, the meta-analyses’ results may be limited by the relatively small number of studies that met the inclusion criteria, as well as their heterogeneity. A Leave-one-out sensitivity analyses for mortality and sepsis showed a decrease in heterogeneity without affecting our meta-analytical results, which supports the robustness and reliability of our findings. A meta-regression could not be done due to the small number of studies.Platelet transfusions in preterm infants are associated to a higher risk of death, sepsis, and NEC and, possibly, to a higher incidence of IVH. Further studies are needed to confirm these associations, namely between PTx and IVH, and to define the threshold from which PTx should be given with less harmful effect. Preferably, RCTs should be conducted in preterm infants, comparing the outcomes of transfused and non-transfused patients at defined and comparable thresholds. Furthermore, it would be valuable to assess outcomes based on the number of transfusions received.Supplementary file1 (DOCX 401 KB)Below is the link to the electronic supplementary material.


Background: Mounting evidence suggests that maternal obesity and gestational weight gain (GWG) may increase the risk of cancer in their offspring; however, results are inconsistent. The purpose of this research is to determine the association between maternal body mass index (BMI) and GWG and the risk of cancer in offspring through a systematic and comprehensive meta-analysis. Methods: A systematic literature search of several databases was conducted on 1 October 2022 to identify relevant studies. The quality of the included studies was evaluated using the Newcastle–Ottawa scale. The overall risk estimates were pooled using a random-effects meta-analysis. Results: Twenty-two studies with more than 8 million participants were included. An increased risk of total cancer was found in offspring whose mothers had a high GWG  but not in offspring whose mothers had a low GWG , when compared with offspring whose mothers had a suitable GWG. In addition, no statistically significant association was found between maternal underweight , overweight/obesity , and risk of total cancer in offspring. Conclusions: Our study proposes evidence that maternal BMI and GWG may be associated with the risk of cancer in offspring, although statistical significance was found only for high GWG. Further well-designed research is required to clarify the potential relevance of maternal BMI and GWG on offspring cancer, especially for specific cancers. As a common cause of death worldwide, cancer brings increasing health and economic burden. It has become a persistent public health challenge and an important obstacle to the increase in human life expectancy ,2. In reStudies have shown that factors such as drinking, smoking, and being overweight/obese contribute to a higher risk of cancer ,5,6. TheSo far, only one relevant meta-analysis published in 2010 focused on the relationship between maternal BMI and risk of testicular cancer in male offspring , in whicTo this end, we aim to perform a systematic and comprehensive meta-analysis regarding the relationship between maternal BMI and the risk of offspring cancer. Our study also quantitatively summarized present evidence on the association between maternal GWG and offspring cancer risk since there was still no relevant meta-analysis. Our meta-analysis may help to access the risk of cancer in offspring associated with maternal factors.This meta-analysis was presented following Preferred Reporting Items for Systematic Reviews and Meta-Analysis (PRISMA) statement and Meta-analysis of Observational Studies in Epidemiology (MOOSE) reporting guidelines ,26. Web 2), normal weight (BMI: 18.5–25 kg/m2), overweight (BMI: 25–30 kg/m2), and obesity (BMI > 30 kg/m2). GWG categories included low GWG (<10 kg or inadequate GWG according to 2009 the Institute of Medicine [IOM] guidelines), normal GWG (10–15 kg or adequate GWG according to 2009 IOM guidelines), and high GWG (>15 kg or excessive GWG according to 2009 IOM guidelines) [The exposure of interest was maternal BMI and GWG. BMI is classified into the following four groups: underweight (BMI < 18.5 kg/mdelines) . OutcomeStudies were considered eligible if they: (1) were published in English; (2) had a case-control or prospective cohort design; (3) had maternal BMI and/or GWG as the exposure was clearly reported; (4) had use of any cancer in offspring; and (5) reported relative risks (RRs), odds ratios (ORs) and hazard ratios (HRs), with corresponding 95% confidence intervals (CIs) to calculate them. The following studies were excluded: (1) letters, case reports, meeting abstracts, or reviews; (2) redundant publications; or (3) studies with unclear or incomplete data. If two or more studies were from the same population, the most comprehensive or latest one was selected.All studies obtained through the search strategies were evaluated by two reviewers (MJX and CY) independently. Any differences of opinion were settled through discussion, and if necessary, the third reviewer (WTT) was invited to have the final vote. Data collection was performed by using a self-made data extraction table to evaluate and extract the following data for each included piece of literature: the first author and year of publication, geographic region, study design, sample size, study population, age of participants, ascertainment of maternal weight, maternal weight categories, outcomes reported, confounds adjusted, and risk estimates with corresponding 95% CIs.The study quality was assessed independently by two reviewers (MJX and CY), with the Newcastle–Ottawa Scale (NOS) for observational studies . The NOSRelative risks (RRs) were used to measure the association between maternal BMI/GWG and cancer in offspring. RRs were considered odds ratios (ORs) because of the low incidence of cancer in offspring . Accordi2 statistic. The Cochran Q test was used to evaluate whether the variation across studies was compatible with chance, and p < 0.1 was considered to indicate significant heterogeneity. The I2 statistic was a quantitative indicator used to evaluate the percentage of the total variance in prevalence estimates due to statistical heterogeneity rather than chance, or sampling error . To explore the sources of heterogeneity, subgroup analyses were conducted based on different categories: maternal BMI , geographic region , study design , study population , ascertainment of maternal weight , and whether confounding factors were adjusted . To assess the robustness of the meta-analysis results, sensitivity analysis was conducted by repeating the meta-analysis after excluding each included study. Begg’s test was used to assess the publication bias. Considering the limited number of included studies, subgroup analyses, sensitivity analysis, and Begg’s test were not performed for specific cancers.All analyses were conducted using R version 4.0.3  and RevMan version 5.3 . The pooled ORs and 95% CIs were computed using a random-effects meta-analysis. The heterogeneity of ORs across studies was assessed using the Cochran Q test and the IA total of 7578 records were identified after retrieval, of which 7466 were excluded through the screening of titles and abstracts. Based on a review of the full texts of 112 studies, 90 studies were excluded, mainly because they were reviews, nondescendant cancers, lack of information on maternal BMI or GWG, unclear or incomplete data, or duplicated data. Finally, twenty-two studies were identified as eligible and included in the present meta-analysis  13,14,1,44,45,46The characteristics of twenty-two studies with 8,329,446 participants are summarized in 2: 0%, p: 0.630). Begg’s test found no potential publication bias . The results of sensitivity analysis suggested that excluding any single research study did not substantially change the overall risk estimates for total cancer, with a range between 1.02 to 1.07 .For maternal underweight, the forest plot of study outcomes is shown in  to 1.07 . Subgrou2: 44%, p: 0.020). Begg’s test found no potential publication bias . Results of the sensitivity analysis suggested that excluding any single research did not substantially change the overall risk estimates for total APOs, with a range between 1.05 to 1.08 .For maternal overweight/obesity, the forest plot of study outcomes is shown in  to 1.08 . SubgrouRisk estimates between maternal overweight/obesity and the risk of specific cancers are summarized in 2: 48%, p: 0.030). Begg’s test found no potential publication bias . The results of the sensitivity analysis suggested that excluding any single research study did not substantially change the overall risk estimates for total APOs, with a range between 1.02 to 1.08 .For low GWG, the forest plot of study outcomes is shown in  to 1.08 . SubgrouRisk estimates between maternal low GWG and risk of specific cancers including leukemia, brain cancer, breast cancer, testicular germ-cell cancer, hepatoblastoma, and neuroblastoma are summarized in 2: 44%, p: 0.040). Begg’s test found no potential publication bias . The results of sensitivity analysis suggested that excluding any single research study did not substantially change the overall risk estimates for total APOs, with a range between 1.08 to 1.12 .For high GWG, the forest plot of study outcomes is shown in  to 1.12 . SubgrouRisk estimates between maternal high GWG and risk of specific cancers including leukemia, brain cancer, breast cancer, testicular germ-cell cancer, hepatoblastoma, and neuroblastoma are summarized in In this meta-analysis, by combining the results of all available cohort and case-control studies with the method of meta-analysis, we provided evidence that maternal high GWG is associated with an increased risk of total cancer with a relative risk estimate of 1.1. Subgroup analysis showed that maternal overweight/obesity was associated with a higher risk of leukemia and a lower risk of testicular germ-cell cancer while no increase in risk was detected for the other subtypes of offspring cancer. As far as we know, this study is the latest comprehensive meta-analysis evaluating the impact of maternal BMI and GWG on the risk of cancer in offspring. Our results may provide valuable and helpful information for women planning pregnancy, pregnant women, and prenatal care providers, and provide another new idea for the primary prevention of cancer.No statistically significant association between maternal BMI and overall cancer in offspring was found in this study. However, when looking at specific cancers, maternal overweight or obesity was found to be a risk factor for leukemia and, interestingly, a protective factor for testicular cancer in offspring. In 2010, a meta-analysis performed by Alam et al. suggested that higher maternal weight did not increase testicular cancer risk in offspring, which is inconsistent with our findings . FurtherWith regard to maternal GWG, a statistically significant association between high GWG and the risk of total cancer in offspring was found. However, when stratified by cancer phenotypes, no significant association was found. Considering the relatively small number of studies included for each specific cancer, the possibility cannot be ruled out that the potential significance of the increased cancer risk cannot be identified due to the limited statistical power. Existing research has shown that insufficient or excessive GWG is not only related to adverse pregnancy outcomes but also has a far-reaching impact on the long-term health of offspring. Excessive GWG is positively associated with a higher risk of C-section, macrocephaly, preeclampsia, gestational diabetes, and postpartum weight retention, while inadequate GWG is also related to a higher risk of placental abruption, small gestational age, preterm birth, and low birth weight ,51,52,53The advantage of our study is the comprehensive search and analysis of all available relevant literature. Before this study, only a meta-analysis published in 2010 assessed the association between maternal prepregnancy BMI and testicular cancer in offspring. Our meta-analysis considered not only maternal BMI, but also the GWG on offspring cancer, and it focused on total cancer and specific cancers. Compared with the previous meta-analysis, our research sample size  is larger, and the analysis is more detailed and comprehensive.One limitation of our meta-analysis is that no source of heterogeneity was found. We performed a subgroup analysis of factors such as geographic region, study design, study population, ascertainment of maternal weight, whether confounding factors were adjusted, and whether GWG was classified according to 2009 IOM guidelines. Unfortunately, no intergroup differences were found. Due to the limited information provided by the original literature, we cannot further discover the sources of heterogeneity. Although most of the included studies adjusted for confounders, these results are not surprising given that cancer is influenced by many factors, and the confounders adjusted for by different studies differ slightly. Another disadvantage is that due to the limited literature available for each cancer, we only analyzed some specific cancers. Future studies need to refine cancer types to clarify the impact of GWG on different cancers. Moreover, although every effort has been made to minimize the possible deviations in the specific retrieval of major databases, there may still be some unidentified literature. Fortunately, as Begg’s test showed, none of our results found publication bias.


Platinum-based chemotherapy is one of the most common treatments for many cancers; however, the effect of chemotherapy varies from individual to individual. Excision repair cross complementation group 1 (ERCC1) is widely recognized as a key gene regulating nucleotide excision repair (NER) and is closely associated with platinum response. Many studies have yielded conflicting results regarding whether ERCC1 polymorphisms can affect the response to platinum and overall survival (OS). Therefore, it is necessary to perform a meta-analysis of patients with specific races and cancer types.Eight databases  were searched. Results were expressed in terms of odds ratios (ORs), hazard ratios (HRs) and 95% CIs.2 = 0%, OR = 6.18, 95% CI, P = 0.003] and ovarian cancer  showed that the rs11615 CT genotype predicted a better response. In the comparison between CC and TT, ovarian cancer  indicated that the CC genotype predicted a better response. In the meta-analysis of OS, the CC genotype was related to longer OS than TT in ovarian cancer .In this study, rs11615, rs2298881 and rs3212986 SNPs were studied. In the comparison between CT and TT on the response to platinum, esophageal cancer [IThe ERCC1 rs11615 polymorphism was related to the response to platinum and OS, but the correlation is based on specific cancer types in the Asian population. Platinum-based chemotherapy is the standard treatment for many cancers, such as ovarian cancer, gastric cancer, esophageal cancer, osteosarcoma, etc. Despite the clear clinical benefit of platinum, some patients still exhibit drug resistance and these patients are increasing with increasing treatment lines. Some studies have shown this phenomenon can be explained by the nucleotide excision repair (NER). DNA damage by platinum is the main anti-cancer mechanism of platinum , 2, and It would certainly benefit patients and clinicians if we could predict the response of patients to platinum before they receive platinum-based chemotherapy. Platinum-sensitive patients receive platinum-containing chemotherapy, while platinum-insensitive patients receive non-platinum chemotherapy or other treatments. It can not only help clinicians to make decisions, but also reduce the pain caused by treatment. Excision repair cross complementation group 1 (ERCC1) component is a mammalian endonuclease that cleaves damaged DNA strands during the repair process of NER and interchain crosslinking. Detection of ERCC1 polymorphisms is expected to predict the efficacy of platinum-based chemotherapy. Two sub-pathways of NER, called global genome repair (GGR) and transcription-coupled repair (TCR), can be mentioned. DNA adducts including those induced by cisplatin are repaired by the nucleotide excision repair (NER) pathway . ERCC1 iMany studies have found conflicting effects of ERCC1 polymorphisms on the platinum response. Gradually, it became clear that the cause of this conflict was ethnic differences. In studies examining the relationship between ERCC1 polymorphisms and the responsiveness of patients with non-small cell lung cancer treated with cisplatin, the results showed that ERCC1 polymorphisms were not associated with platinum responsiveness and there was no evidence of a potential prognostic role in a Caucasian population , 7. WhilCurrently, the most studied SNPs are rs11615, rs2298881 and rs3212986. Because the current study showed that the results of ERCC1 SNPs as a predictor of platinum treatment response and prognosis were contradictory, we performed a systemic review and meta-analysis to assess the evidence of effects of ERCC1 rs11615 C>T, rs2298881C>A and rs3212986 C>A SNPs on the efficacy of platinum-based chemotherapy and overall survival.The systematic review was performed according to the Meta-Analysis of Observational Studies in Epidemiology , Preferred Reporting Items for Systematic Reviews and Meta-analysis (PRISMA) statement and protocol. The systematic review protocol was registered with the international prospective register of systematic reviews (PROSPERO): CRD42022339301.Two reviewers (CH J and ZL Z) independently searched databases including EMBASE, PubMed, Cochrane Library, Chinese National Knowledge Infrastructure (CNKI), Scopus, VIP, China Biology Medicine disc (CBM) and Wanfang from inception to June 2022. The search strategy is displayed in the Inclusion criteria: (1) Population: Asian patients who were included in detection (primary detection) or being followed for recurrent disease and had cancer recurrence detected in histopathology. Patients who were receiving chemotherapy including cisplatin, carboplatin, oxaliplatin or nedaplatin. (2) Intervention: ERCC1 polymorphisms, including rs11615 (CT vs CC vs TT), rs3212986 (AA vs CA vs CC) and rs2298881 (AA vs CA vs CC). (3) Control: each phenotypic data was served as a comparator with the other two . (4) Outcome: Primary outcome is served as the response to platinum-based chemotherapy in each SNP. After receiving platinum containing chemotherapy, patients with CR or PR were defined as having good response, while those with SD or PD were defined as having poor response. The secondary outcome was defined as OS for each SNP. OS was defined as the time from the observation point to death from any cause. (5) Study design: cohort and case control studies.Exclusion criteria: All studies with mutated bases in gene fragments that did not meet the requirements listed in the inclusion criteria were excluded. We also excluded studies reporting data on laboratory developments tests as well as reviews, letters to editors, editorials, study protocols, case reports, brief correspondence, and articles published in languages other than English or Chinese. The references of all papers included were scanned for additional studies of interest.Three investigators independently extracted the following information from the included articles: author name, publication year, country, chemotherapy regimen, number of patients, cancer species, number of good responses (CR+PR) and poor response (SD+PD), hazard ratio and 95% confidence interval between different genotypes and overall survival. The Newcastle‒Ottawa Quality Assessment Scale was used to assess the quality of each of the included studies. Any discrepancy was resolved by discussion or by involving an arbiter (XQ W).We assessed the risk of false positives or false negatives by TSA in a meta-analysis. Sequential monitoring boundaries were established to limit Type I error to 5% and Type II error to 20% to detect responsiveness to platinum and OS of patients under various models. When cumulative Z curves crossed conventional boundaries and trial sequential monitoring boundaries and the sample size reached the accrued information size (AIS), the expected intervention effect may reach a sufficient level of evidence. TSA was completed using the Stata, version 13.0, R package "foreign" and "ldBounds".2; I2 greater than 50% and a P value less than 0.10 suggested significant heterogeneity, and a random effect model was applied; I2 less than 50% and a P value more than 0.10 suggested tolerable heterogeneity, and a fixed effect model was applied. Sensitivity analysis was performed by sequentially omitting each study to examine the robustness of the results. Potential publication bias was evaluated using a funnel plot (number of studies ≥ 10) and Begg’s and Egger’s tests. If significant publication bias existed, the trim and fill method was performed to validate the robustness of the meta-analysis results. Sensitivity analysis adopts the row-by-row division method. All statistical analyses were calculated via Stata, version 13.0 . P values < 0.05 and Bonferroni-corrected p < 0.017 in studies with the response to platinum as the outcome (Results 2) and p values <0.05 and Bonferroni-corrected p < 0.025 in studies with OS as the outcome (Results 3), were defined as statistically significant, except those for heterogeneity.The effect estimates of interest were the OR (95%CI) in the meta-analysis of response to platinum and HR (95%CI) in the meta-analysis of overall survival. Studies exploring the response to platinum and overall survival of cancers were analyzed separately, and subgroup analyses were performed according to different cancer species. Assessment of heterogeneity was performed using Higgins IWe identified 42 studies in CBM, 228 studies in Embase, 274 studies in the PubMed database, 32 studies in the Cochrane Library, 218 studies in the CNKI database, 86 studies in the VIP database, 20 studies in Scopus, and 306 studies in the Wanfang database. The flowchart of study selection is shown in 2.1.1 Dominant model (CT+TT vs CC). The rs11615 polymorphism has a wild-type C allele and a variant T allele. A total of 26 studies  and ovarian cancer  showed that the rs11615 CT genotype predicted a better response to platinum. In gastric cancer , lung cancer , colon cancer , nasopharyngeal , cervical cancer  and osteosarcoma , CT genotype seems to be not related to the response to platinum. The forest plot is shown in The funnel plot was basically symmetrical (2.1.3 Homozygous model (CC vs TT). Number of studies and people involved in the study as described in 2.1.1. ERCC1 rs11615 CC genotype was related to a good response to platinum-based chemotherapy. Homozygous model (CC vs TT): I2 = 53.6%, random effect model was used, OR = 1.84, 95%CI, P<0.001, as shown in 2 = 0%, OR = 3.85, 95%CI , P = 0.021], ovarian cancer  and colon cancer  showed that CC predicted a better response to platinum than the TT genotype. In gastric cancer , lung cancer , nasopharyngeal , cervical cancer  and osteosarcoma , CC vs TT seems to be not related to the response to platinum. The subgroup plot is shown in The funnel plot was basically symmetrical (2.2.1 Dominant model (CA+AA vs. CC). The rs2298881 polymorphism has a wild-type C allele and variant A allele. A total of 6 studies , osteosarcoma  and ovarian cancer  groups, as shown in  studies , 37, 38,The funnel plot was basically symmetrical (2.2.2 Heterozygous model (CA vs CC). Number of studies and patients involved in the study as described in 2.2.1. The CA and CC genotypes showed no difference in platinum sensitivity. Heterozygous model (CA vs CC): I2 = 45.9%, fixed effect model was used, OR = 0.94, 95% CI, P = 0.653, as shown in 2 = 42.2%, OR = 0.75, 95% CI , P = 0.495], osteosarcoma  and ovarian cancer  groups, as shown in The funnel plot was basically symmetrical (2.2.3 Homozygous model (AA vs CC). Number of studies and patients involved in the study as described in 2.2.1. The AA and CC genotypes showed no difference in platinum sensitivity. Homozygous model (AA vs CC): I2 = 76.2%, random effect model was used, OR = 0.90, 95% CI, P = 0.758, as shown in 2 = 45.8%, OR = 0.69, 95% CI , P = 0.248], osteosarcoma  and ovarian cancer  groups, as shown in The funnel plot was basically symmetrical (2.3.1 Dominant model (CA+AA vs. CC). The rs3212986 polymorphism has a wild-type C allele and variant A allele. A total of 9 studies , osteosarcoma  and ovarian cancer  groups. The CA+AA vs. CC genotype was related to the response to platinum in lung cancer . The results are shown in  studies , 33–36, The funnel plot was basically symmetrical (2.3.2 Heterozygous model (CA vs CC). Number of studies and patients involved in the study as described in 2.3.1. The CA and CC genotypes showed no difference in response to platinum. Heterozygous model (CA vs CC): I2 = 55.8%, random effect model was used, OR = 0.91, 95% CI, P = 0.551, as shown in 2 = 43.6%, OR = 1.04, 95% CI , P = 0.821], osteosarcoma  or ovarian cancer . The CA vs CC genotype was related to the response to platinum in the lung cancer  groups, as shown in The funnel plot was basically symmetrical (2.3.3 Homozygous model (AA vs CC). Number of studies and number of patients involved in the study as described in 2.3.1. The AA and CC genotypes showed no difference in response to platinum. Homozygous model (AA vs CC): I2 = 66.1%, random effect model was used, OR = 0.96, 95% CI, P = 0.883, as shown in 2 = 56.2%, OR = 1.19, 95% CI , P = 0.562], osteosarcoma  and lung cancer  groups. The AA vs CC genotype was related to the response to platinum in the ovarian cancer  group. The above results are shown in The funnel plot was basically symmetrical (3.1.1 Homozygous model (TT vs CC). A total of 8 studies  and osteosarcoma  groups. The CC genotype was related to longer overall survival in ovarian cancer , as shown in  studies , 24, 34 Begg’s test and Egger’s test showed that P = 0.902 and 0.530, respectively, indicating that there was no publication bias. The sensitivity analysis showed that the result was reliable (3.1.2 Heterozygous model (CT vs CC). A total of 7 studies , gastric cancer  and ovarian cancer  groups. The forest plot is shown in  studies , 24, 34 Begg’s test and Egger’s test showed that P = 0.548 and 0.398, respectively, indicating that there was no publication bias. The sensitivity analysis showed that the results were not robust (3.2.1 Homozygous model (AA vs CC). A total of 3 studies [2 = 26.4%, fixed effect model was used, HR = 1.80, 95% CI, P = 0.044, as shown in  studies , 20, 38 3.2.2 Heterozygous model (AC vs CC). Number of studies and number of patients involved in the study as described in 3.2.1. AC vs CC was not related to overall survival. Heterozygous model (AC vs CC): I2 = 0%, fixed effect model was used, HR = 1.17, 95% CI , P = 0.416, as shown in 3.3.1 Heterozygous model (AC vs CC). A total of 5 studies [2 = 0%, fixed effect model was used, HR = 1.24, 95% CI , P = 0.108, as shown in  studies , 34–36 iBegg’s test and Egger’s test showed P = 1.000 and 0.580, respectively, indicating that there was no publication bias. The sensitivity analysis showed that the result was reliable (3.3.2 Homozygous model (AA vs CC). Number of studies and number of patients involved in the study as described in 3.3.1. Homozygous model (AA vs CC): I2 = 59.4%, random effect model was used, HR = 1.82, 95% CI, P = 0.109, as shown in Begg’s test and Egger’s test showed that P = 0.806 and 0.422, respectively, indicating that there was no publication bias. The sensitivity analysis showed that the result was reliable  between rs11615 SNP and other nearby alleles that may cause platinum resistance . The difOur study not only analyzed the platinum sensitivity and prognosis of different SNPs but also compared different cancers horizontally. It can be seen that different cancers and different SNPs have varying effects on platinum sensitivity and prognosis. Ovarian cancer identified in this study was closely related to ERCC1 gene polymorphisms. Ovarian cancer patients do benefit from DNA damage repair pathway inhibitors, such as Rucaparib –49, OlapThe ERCC1 rs11615 CT and CC genotypes predicted a better response to platinum than the TT genotype in esophageal and ovarian. The CC genotype was related to longer overall survival than TT in ovarian cancer. The ERCC1 rs2298881 and rs3212986 A/C polymorphisms were not related to the response to platinum and OS.S1 Checklist(DOCX)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.


In summary, the rs231775 locus polymorphism in the CLTA-4 gene may be a risk factor for GD in Asian children. The G allele may be a susceptibility factor, while the allele A may be a protective factor against GD in Asian children. In the future, more large-scale studies may be needed to verify our results.The aim of this study was to systematically evaluate the correlation between the rs231775 locus polymorphism in the cytotoxic T lymphocyte-associated antigen 4 (CTLA-4) gene and genetic susceptibility to Graves’ disease (GD) in children. Some studies found that the CTLA-4 gene polymorphism was associated with GD in children. The data up to February 2022 were retrieved from the databases. Stata 15.0 software was used for meta-analysis. A total of seven studies were included in our research. The results of the meta-analysis showed that the rs231775 locus polymorphism in the CTLA-4 gene in general and Asian populations was correlated with children’s susceptibility to GD ; GG vs AA: OR = 1.34, 95% CI (1.04–1.73); AG vs AA: OR = 1.32, 95% CI (1.02–1.10); AG + GG vs AA: OR = 3.81, 95% CI (2.17–6.70); GG vs AA + AG: OR  The development of GD is thought to be determined by genetic and environmental factors, and in terms of genetic factors, it has shown a gradual increase in recent years. Also, it is becoming a common disease ,2. It th22.1We searched PubMed, Embase CBM, Cochrane Library, cnki.net, Wanfang Data, and VIP databases with words such as “CTLA-4, gene polymorphism, hyperthyroidism, GD, children, and correlation.” The search period included data from the time the database was established to February 2022. The databases were searched for free words and subject words. After multiple pre-retrieval, the search strategy was confirmed.2.2https://www.crd.york.ac.uk/prospero/#recordDetails).The present systematic review and meta-analysis were carried out according to the established protocol (PROSPERO: CRD42022343849) . (Available from: 2.3(1) Study on the relationship between cytotoxic T lymphocyte-associated antigen-4 (CTLA-4) gene polymorphism and genetic susceptibility (GD) to hyperthyroidism (limited to Chinese and English languages); (2) Patients with a confirmed diagnosis of GD and under 18 years of age; (3) The genotype frequency of the control group determined according to the Hardy–Weinberg equilibrium.2.4in vitro cell experiments; (2) Study on the relationship between gene polymorphism at a different locus and GD; (3) Non-case-control study; (4) Incomplete original data; (5) Studies not having direct access to patient age, sample size, and study data.(1) Study including review, animal experiments, conference papers, and 2.5Two staff members performed data extraction and quality evaluation along with further cross-verification. A third researcher was also added to the discussion, who solved the issues in case of differences. After the literature selection, the extracted data included the first author, publication year, sample size, age, and specific distribution of the genotypes  in both the case and control groups. Hardy–Weinberg (H–W) genetic balance test was performed to distribute genotypes between the control groups.2.6As this meta-analysis included only case-control studies, the quality of the literature was scored according to the Newcastle-Ottawa Scale (NOS), which evaluated case-control studies by three blocks with eight items, including study population selection, comparability, and exposure assessment. The NOS score ranged from 0 to 9, with NOS score ≥ 7 as high quality and NOS score < 7 as low quality. Two reviewers evaluated the NOS score to assess the risk of bias in the included studies. In case of disagreement, it was resolved by discussion or by a third party .2.7P < 0.05, while the heterogeneity was quantitatively evaluated using the I2 value. The result of the heterogeneity test at P < 0.05 or I2 > 50% indicated a great heterogeneity among the studies. Thus, the heterogeneity hypothesis should be explored in a series of sensitivity analyses adopting a random effect model. If this strategy is not used, a fixed-effect model may be applied. (3) The Mantel–Haenszel (M–H) method was used to calculate the combined OR value and its 95% CI. A 95% CI value of less than 1 indicated a significant difference between the case and control groups. (4) The inverted funnel diagram and Begg’s and Egger’s tests were used to determine any publication bias. P ≥ 0.05 indicated that there was no publication bias. Furthermore, the trim-and-fill method was used to test and correct its possibility. A 95% CI value of less than 1 indicated that there was no publication bias.(1) The H–W equilibrium test was adopted to distribute genotypes in the study. (2) Stata 15.0 software was applied for the meta-analysis. Heterogeneity between studies was evaluated using the Cochran Q test. The obvious heterogeneity was found in the data from the published literature with 33.1According to the NOS evaluation scale, the literature scores were above 7, which could be included in this systematic review. The details are displayed in 3.2P ≥ 0.05,  that met the inclusion criteria were selected for meta-analysis. The included literature belonged to mainland China, Hongkong, Japan, Poland, Brazil, and five other countries and regions. In each study, the genes in the control group followed the H–W equilibrium, while the test values of each group were found to be 3.3I2 < 50%, P > 0.05), the fixed effect model was adopted. The dominant model was AG + GG vs AA. Since statistically significant differences were observed in the heterogeneity test among studies , the random effect model was applied, as shown in The allele model used was A vs G; co-dominant model: GG vs AA and GA vs AA; recessive genetic model: GG vs AG + AA; and super dominant model: GG + AA vs AG. Since the heterogeneity test among the studies did not show a statistically significant difference : OR = 0.75, 95% CI (0.66–0.86); co-dominant model (GG vs AA): OR = 1.34, 95% CI (1.04–1.73); co-dominant model (AG vs AA): OR = 1.32, 95% CI (1.02–1.10); dominant model (AG + GG vs AA): OR = 3.81, 95% CI (2.17–6.70); and recessive model (GG vs AA + AG): OR = 1.23, 95% CI (1.05–1.45). The results are shown in The results of the analysis suggested that the G allele significantly increased the risk of GD in the Asian population and the total population, which was analyzed by comparing the G and A alleles of the CTLA-4 + 49 G allele, as shown in Additionally, no correlation was found between GD susceptibility and CTLA-4 + 49 locus polymorphism under the super dominant model AA + GG vs AG , which is shown in 3.4Yanagawa et al.  found a 3.5P values of Begg’s and Egger’s test results were between 0.230 and 0.848, with P ≥ 0.05. The trim and fill analyses showed that the OR value and 95% CI of each gene model showed no change before and after this method. As shown in The 4The decrease in CTLA-4 expression on the cell surface and its functional stability is believed to increase T-cell activation, which can lead to the development of autoimmune diseases. In the past 10 years, studies at home and abroad have shown that  CTLA-4 iP < 0.01) in the entire population for the G allele and the genotypes GG, AG, and AG + GG of the CTLA-4 gene locus rs231775. The subgroup analysis, based on regions, showed that the frequency of the G allele and the genotypes GG, AG, and AG + GG in the case group was significantly higher than in the control group, with statistically significant results (P < 0.01). The rs231775 locus was not associated with GD in children of Caucasian populations, which proved regional differences between the polymorphism of the rs231775 locus of the CTLA-4 gene and the susceptibility to GD in children. The above results showed that the CTLA-4 + 49 A allele could reduce the risk of GD. Compared to the AA genotype, GG and AG genotypes might show an increased risk of GD in children. The G allele may be the risk factor for GD in children, while the A allele may be the protective factor.To perform the systematic meta-analysis, we selected case-control study exploring the correlation between CTLA-4 + 49 A/G polymorphism and GD in children belonging to different regions. The difference was found to be statistically significant between the case and control groups . However, this locus is not related to GD in Caucasian children. One reason for this could be the inclusion of fewer Caucasian population studies in the literature, while another could be ethnic heterogeneity. Therefore, to avoid covering up its real effect, different populations should not be blindly combined when performing research and analysis. Also, this study has some limitations. First of all, meta-analysis is a descriptive secondary analysis that is based on published studies. The results of the meta-analysis may be affected by different genetic testing methods. Only a few studies were included. Different studies addressed heterogeneity, the study design of variables, and different statistical methods , which mP < 0.05 means small study effects. (2) Begg’s test is also known as the rank correlation test. Its main idea is to test the correlation between standardized effects and effect variance based on Kendall’s tau rank correlation test, that is, to test the correlation between effects and sample size. Under the null hypothesis of no publication bias, standardized effects can be considered independent and identically distributed, and there is no correlation between standardized effects. The test results depend primarily on the p-value, and publication bias is generally believed to exist if P < 0.05. (3) Based on the hypothesis that publication bias causes funnel plot asymmetry, the cut-and-complement method uses the iterative method to estimate the number of missing studies, but its significance is not to estimate the specific number of missing studies. After adding a part of the studies, a meta-analysis will be conducted again. If the estimated value of the combined effect size does not change significantly, it indicates that publication bias has little influence, and the result is relatively robust. In this study, the P values of Begg’s and Egger’s test results were between 0.230 and 0.848, P ≥ 0.05. The results of the shear and complement analysis showed that the OR values and the 95% CI of each gene model were basically unchanged before and after the shear and complement analysis, and the conclusions were consistent, indicating that the differences were not statistically significant [The interpretation of publication bias: (1) Egger’s test, also known as the linear regression method, is the normal standard deviation of a study included in the meta-analysis. The regression equation SND = A + Bx precision was established for SND and precision. If there is no publication bias, the intercept (A) of the regression line is 0. If the regression line does not pass through the origin, then the larger the intercept A, the higher the degree of bias. The detection result nificant ,21.Existing studies have shown that the CTLA-4 gene polymorphism is closely correlated with complications, drug effect, and GD recurrence . The RCT


I2, as well as several subgroups, meta-regression and sensitivity analysis were performed to identify the source of heterogeneity. Overall, 17 eligible studies containing 6285 cases and 13 909 healthy controls were included. The REM showed a significant positive association between malaria infection and epilepsy development . In subgroup analyses, significant positive associations were observed in studies that: epilepsy was the outcome in the follow-up of patients with cerebral malaria ; used blood smear to diagnose malaria ; included only children ; published before 2010 . Our findings indicated that patients with malaria, especially those with cerebral malaria, are at a high risk of epilepsy development; however, further well-designed and controlled studies are needed to verify the strength of the association.Epilepsy, a chronic disease of the central nervous system, is highly prevalent in malaria-endemic regions. Therefore, several studies have evaluated the associations between malaria infection and epilepsy development. A meta-analysis of observational studies published from inception to 10 May 2022 has been conducted to synthesize and pool the existing data on this topic. The relevant publications were systematically searched in PubMed/Medline, Scopus, Embase and Web of Science database collections. A random-effects meta-analysis model (REM) was utilized to generate the pooled odds ratio (OR) at 95% confidence intervals (CIs). The between-studies heterogeneity was assessed with  EpilepsPlasmodium parasites, transmitted to humans by the bite of the infected female Anopheles mosquito  and P. falciparum (dominant in Africa) are the most common species with the greatest threats  statement as guidance to design the study and report and interpret the findings . All retrieved publications were exported to EndNote Reference Manager X8  where duplicates were removed.Four literature databases were systematically searched for relevant publications, including PubMed/Medline, Scopus, Embase and Web of Science collection, from inception to 10 May 2022. The inclusive keywords composed of malaria, severe malaria, cerebral malaria, brain malaria, After duplicate removal, 2 reviewers (A. A. K. and A. R.) independently screened the titles and abstracts of identified references, followed by full-text screening based on predefined inclusion and exclusion criteria. All the discrepancies were resolved by consulting with the lead investigator (A. R.). The main criteria for inclusion were: (1) peer-reviewed observational studies with cross-sectional, cohort or case–control design; (2) studies that included patients with confirmed malaria infection or patients with epilepsy as the cases, and the controls were individuals without malaria, or without the epilepsy/neurological complications; (3) studies that used internationally recognized diagnostic assays or guidelines to diagnose malaria infection or epilepsy; (4) studies that the risk point estimate was reported as an odds ratio (OR) and confidence intervals (CIs), or the information was presented such that an OR and 95% CI could be calculated. The exclusion criteria for this study included (1) studies that failed to quantitatively assess the relationship between malaria and epilepsy; (2) studies that an OR and 95% CI could not be calculated; (3) studies without original data (e.g. systematic reviews and letters); (4) case reports and case-series studies; (5) full-text of the article was not available (e.g. abstract or conference articles).et al., et al., Two reviewers (A. T. and A. A. K.) independently extracted the following information from each eligible study using a standardized data extraction form: first author, publication year, type of participants , study design , type of population , country, diagnostic assay for malaria infection , the number of cases and control subjects, the prevalence of epilepsy or malaria in individuals of each of the subject groups. The quality of cohort and case–control studies was assessed by the Newcastle–Ottawa Scale (NOS) (Text S1), as endorsed by the Cochrane network  were considered statistically significant.Meta-analyses were performed using Stata software ver. 17 . The pooled prevalence of epilepsy or malaria infection for each case and control group was estimated at a 95% CI using the DerSimonian–Laird random-effects model (REM)  and case–control (n = 8) study designs, while all malaria-based studies had cohort study designs (4 prospective and 2 retrospective). Participants in 12 and 3 datasets were only children and only adults, respectively, while in 9 datasets, both children and adults participated. Considering diagnostic methods, 11 datasets used serological methods, 11 used blood smear parasitology methods and 2 used medical history for severe malaria. According to NOS, the overall quality of datasets was high for half of the datasets (n = 12) and moderate for the other half , it was low or moderate in some sub-groups, especially for those indicating significant positive association, including malaria-based studies , retrospective , prospective cohorts  and studies published before 2010   could be the source of heterogeneity and accounted for 64.5% of between-study heterogeneity.Although the overall heterogeneity was high (= 0.184) . MultivaThe sensitivity analysis was conducted to evaluate the possible influence of any individual study on the main results. The analysis assessed whether omitting 1 study substantially altered the main outcome or magnitude of the summary estimates of the remainders. This analysis indicated that the exclusion of any individual study did not significantly alter the overall results of the meta-analysis , implyinA cumulative meta-analysis was also performed to evaluate the consistency of the evidence over the years as recent studies were added. The cumulative analysis indicated that between 2003 and 2022, there was a positive association between malaria infection and epilepsy development but with a swinging effect size and narrowing CIs .Fig. 4.Epilepsy is a highly prevalent neurological disorder in LMIC, especially in malaria-endemic areas; however, a definitive causative relationship is yet to be established. The results of this systematic review and meta-analysis are consistent with a previous study (Christensen and Eslick, et al., Plasmodium berghei infection in C57BL/6J mice (Oca et al., et al., et al., et al., et al., et al., et al., et al., et al., The association between malaria and epilepsy development is likely explained by more than 1 mechanism, possibly due to CM and its related inflammation (Singh Although the present study comprehensively evaluated the associations between malaria infection and epilepsy, some limitations should be acknowledged in the interpretation of our findings. First, some studies used medical history or serological methods to define malaria, which could be subject to diagnosis biases in pooled OR; however, subgroup analysis was performed in this study according to diagnostic criteria to overcome this limitation. Second, there were insufficient data considering sex, age or other comorbidities, and notably, information about other infections was underrepresented in some included studies. Despite these drawbacks, the findings and interpretations presented here provided useful insights concerning the association between malaria infection and epilepsy.In conclusion, our findings indicated that patients with malaria, especially those with CM, are at a higher risk of epilepsy development; however, several gaps remain in our understanding of pathophysiological mechanisms of this event. Therefore, further human or experimental studies are needed to focus on long-term sequelae of CM such as epilepsy; neuropathological and physiological changes leading to these sequelae; the impact of therapeutic agents that are used during CM episodes; and the impact of other epileptogenic intracranial infections on the development of epilepsy and other neurological sequelae.


Despite decades of research, causes of ALS remain unclear. To evaluate recent hypotheses of plausible environmental factors, the aim of this study was to synthesize and appraise literature on the potential associations between the surrounding environment, including urbanization, air pollution and water pollution, and ALS.We conducted a series (n = 3) of systematic reviews in PubMed and Scopus to identify epidemiological studies assessing relationships between urbanization, air pollution and water pollution with the development of ALS.The combined search strategy led to the inclusion of 44 articles pertaining to at least one exposure of interest. Of the 25 included urbanization studies, four of nine studies on living in rural areas and three of seven studies on living in more highly urbanized/dense areas found positive associations to ALS. There were also three of five studies for exposure to electromagnetic fields and/or proximity to powerlines that found positive associations to ALS. Three case-control studies for each of diesel exhaust and nitrogen dioxide found positive associations with the development of ALS, with the latter showing a dose-response in one study. Three studies for each of high selenium content in drinking water and proximity to lakes prone to cyanobacterial blooms also found positive associations to ALS.Whereas markers of air and water pollution appear as potential risk factors for ALS, results are mixed for the role of urbanization. Amyotrophic lateral sclerosis (ALS) is a rare neurological disease characterized by selective loss of upper and lower motor neurons that ultimately results in the progressive loss of voluntary muscle function , 2. The Because the ≥658 mutations in 126 genes identified to be linked to ALS only explain a small fraction of all ALS cases , 15, thePrevious studies on the presence of environmental factors in proximity to one's residence also point to a possible relation to ALS. For example, a case-control study found that agricultural activity in the region, rather than rural residence itself, was significantly associated to ALS . Furthern = 3). Indexed articles since database inception were searched in PubMed and Scopus using a combination of keywords/MeSH terminology and citation analysis .Cases of ALS are identified with ICD-8 code 348.0, ICD-9 code 335.20, or ICD-10 code G12.21 or through medical records, or are diagnosed by a physiatrist, neurologist, and/or specialist using electrodiagnostic or clinical diagnosis criteria  were alInformation on authors, year of publication, study region, study design, study participants, ALS diagnostic criteria, source of cases/exposed, source of controls/unexposed, environmental risk factors of interest, method of exposure ascertainment, and main findings pertaining to environmental risk factors of interest were extracted from the included articles by author DS. Extraction of information for articles excluded after full-text screening was limited to authors, year of publication, study region, title, and the main reason for exclusion. Included articles had their quality assessed by author DS using either the Newcastle-Ottawa Scale   or a ModThe combined search strategy focused on identifying epidemiological studies that related to urbanization, air pollution and water pollution in the development of ALS and produced 1,388, 434 and 1,215 articles for each exposure of interest, respectively, after removal of duplicate citations –C. Thesevia structured interviews  and controls  and utilized geospatial data  had an elevated standardized incidence ratio (SIR) of ALS of 1.28 , as well as communes with waste disposal activities and companies involved in cleaning and/or discharge , while those communes with extractive industries, storage of pollutants industries or water treatment plants did not have an elevated SIR . In contOnce adjusted for various potential confounders, there were no significant difference between the individuals with ALS and controls for distance to the nearest petrol stations, residential streets, and green areas in a nested case-control study by Povedano et al. in Spain .Malek et al.  were the2.5) absorbance , nitrogen dioxide , and nitrogen oxides (NOX) concentrations  in the fully adjusted models. PM2.5 absorbance and NO2 remained significant after adjusting additionally for urbanization degree and showed linear trends . Such findings were also observed in a follow-up research letter to this Dutch study (2.5 and particulate matter (PM10) elemental fractions of road traffic non-tail pipe emissions, which were associated with significantly higher ORs for ALS in copper, iron, nickel, sulfur, silicon (Si) and vanadium particulate and fine particulate elemental constituents. However, only the independent associations between NO2, PM10Si, and the odds of ALS remained after correction for NO2 in two-pollutant models. Ultrafine particles were not associated with ALS. In both the original study and the follow-up research letter, PM10 itself was not associated with the odds of ALS , hydrogen sulfide and nitrogen oxide (NO) were also not associated with ALS. Statistically significant differences between the cases and the controls were seen, however, for exposure levels of some atmospheric contaminants [sulfur dioxide [SO2], ozone [O3] and benzene], metals  and in the levels of arsenic  of ALS: Quartile 2 , quartile 3 , and quartile 4 .A case-control study in Spain also reported no association between PM arsenic . However2 was not only significantly associated with increased odds of ALS in all three papers that studied this pollutant  in the fully adjusted logistic regression model. However, a recent Danish case-control study by Bellavia et al. found no significant association between diesel exhaust and ALS   .p = 0.03) and silica (p = 0.01) were significantly associated to ALS on a continuous scale. However, only silica had a significant OR for both low  and high  exposures relative to those never exposed. Other occupational particulates investigated included animal contact and endotoxin, as well as some hazardous air pollutants (asbestos and polycyclic aromatic hydrocarbons), none of which were significantly associated with ALS on a continuous scale.The case-control study by Visser et al. studied occupational exposure to other work-related air pollutants, particularly various particulates, through employment of job exposure matrices alongside face-to-face interviews and structured questionnaires . OccupatOf the 3 studies that assessed exposure to high selenium content in water as a risk factor within a common study population, all of them found it to be positively associated with ALS. The residents of Reggio Emilia, Italy, were accidentally exposed to drinking water with high selenium content until 1988 with initial exposure starting potentially as early as 1972 . The expOutside of Reggio Emilia, another cohort study, a retrospective follow-up between 1979 and 2008, compared mortality rates between 154,932 marine and naval personnel exposed to drinking water contaminated by solvents (Camp Lejeune) and 154,969 marine and naval personnel unexposed to contaminated drinking water (Camp Pendleton) . Water sThere was one other paper that studied contaminated water . The stuBeyond contaminated water, there were three papers identified that studied source of drinking water as a risk factor for ALS , 63, 65 Of the three studies that assessed proximity to water bodies as a risk factor, two found it to be positively associated with ALS. A case-control study in the New England region of the United States of America by Andrew et al. indicated that residing full-time within 2 miles of a water body  is positively related to ALS status . HoweverThree studies all within the Northern New England region identified a potential link between lakes prone to algae blooms and ALS , 69, 70.Collectively, water-quality indicators and lake conditions that promote harmful algal blooms seem to be associated with ALS in the Northern New England region and should be studied further in epidemiological study designs higher up on the evidence hierarchy, as well as in other regions with lakes prone to cyanobacterial blooms.In addition to inquiring about waterbody proximity among residents of New England, the case-control study by Andrew et al. in the United States of America included questions pertaining to participation in water sports in their questionnaire . AlthougThis series of three systematic reviews identified 44 epidemiological studies assessing relationships between urbanization, air pollution and water pollution with the development of ALS. Many potential environmental risk factors were identified that are consistently associated with a higher risk of ALS, capturing the ever-growing literature on environmental factors since previous systematic reviews. This is the first study to systematically synthesize all of this for three different types of environmental exposures  and expands on previous systematic reviews that pertained to rural environments  or varioWhile few studies found an association between rural areas and ALS , 44, 46,Exposure to electromagnetic fields and high-voltage infrastructure in the surrounding area showed an increased to the risk of ALS in three of five studies , 42, 51,As for the various industrial related activities studied in a limited number of included articles with mixed results, future research should aim to use geospatial data as opposed to self-administered questionnaires in case-control study designs when databases are available, such as the geodata and maps used in the various electromagnetic field studies , 42, 51.2 was significantly associated with increased risk of ALS in all three papers that studied this pollutant . Among the included studies pertaining to diesel exhaust exposure, three found a significant positive association with ALS  exist and should be consulted for those seeking further information on their potential role in ALS etiology  were not conducted in the three included studies , 62, 64.For water sports, only frequent participation in water-skiing was significantly associated with ALS in the final adjusted models . The stuThe hypothesis in the field of ALS that the disease is a complex genetic condition means that contributions from multiple genetic and environmental risk factors are likely involved in the development of ALS, with relative effect size from individual factors being rather small . EnvironThe use of a combination of keywords/MeSH terminology to identify records was a strength for this series of systematic reviews . The stuThe current series of three systematic reviews sought to appraise the growing literature of environmental factors pertaining to urbanization, air pollution and water pollution in the development of ALS. Studies on environmental factors pertaining to urbanization of the surrounding area were the most studied in the context of this series of systematic reviews but presented mixed results for an association to ALS. The relationship of air pollutants to ALS was clearer with results suggesting that diesel exhaust exposure and nitrogen dioxide, a primary product of diesel combustion, are linked to an increased risk of ALS. Heavy metal contamination in water, particular selenium, as well as proximity of residence to lakes prone to cyanobacterial blooms, was associated with an increased risk of ALS. Measures to reduce air and water pollution are important as diesel exhaust, nitrogen dioxide and poor water quality all appear to represent modifiable risk factors of ALS.The original contributions presented in the study are included in the article/DS conceived the current study, performed the literature search, appraised the included articles, extracted the pertinent data, and drafted the initial manuscript. PPWR independently appraised and extracted data from a subsample of the included articles. MB resolved any disagreements between DS and PPWR. DS, MB, and CO'C revised the intellectual content of the work critically. All authors approved the final version to be published.