Discovery and validation of genetic factors for multifactorial and polygenic disorders like stroke are needed to make progress in precision medicine. Although some traditional risk factors for stroke have been identified, they do not fully explain the pathophysiological mechanism of ischemic stroke. The research of genetic risk factors is becoming increasingly relevant in the understanding of stroke mechanisms and the finding of population-specific therapeutic targets. The methylenetetrahydrofolate reductase (MTHFR) gene is involved in homocysteine metabolism, and a high homocysteine level is a risk factor for stroke. Using a meta-analysis technique, we investigated the link between the MTHFR C677T gene polymorphism and the risk of ischemic stroke.We used the electronic databases PubMed, Medline, Embase, and Google Scholar to find articles in the Journal of Stroke. If heterogeneity was more than 50%, pooled odds ratios (ORs) with 95% confidence intervals (CIs) were calculated using a random-effects model; otherwise, a fixed-effects model was used.A total of 67 case-control studies with 17,704 cases and 21,981 controls met our inclusion criteria. The Asian population was represented by 41 studies, whereas the Caucasian population was represented by 26. Under the recessive model, a gene polymorphism at the 677 location of the MTHFR gene is related to an elevated risk of ischemic stroke .People who have the MTHFR C677T gene polymorphism have a greater risk of stroke than people who do not. Stroke has risen to become the second largest cause of mortality in adults and the third leading cause of disability. Understanding the pathogenesis of stroke necessitates the finding of risk factors ,2. TradiThe 5,10-methylenetetrahydrofolate reductase is an important enzyme that regulates the metabolism of homocysteine (Hcy) levels . MTHFR iMethodology and literature searchWe followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines for reporting meta-analysis findings . We condInclusion and exclusion criteriaInclusion criteria included the following: (a) studies that used a case-control study design investigating the relationship between the MTHFR C677T gene and the risk of ischemic stroke; (b) studies including ischemic stroke cases and healthy controls; (c) studies that mentioned the diagnostic criteria for ischemic stroke; (d) studies that reported the genotypic frequencies for both cases and controls; (e) studies with patients aged > 18 years; and (f) studies with enough data for extraction for computing pooled effect size.Studies were excluded (a) in case genotype frequencies could not be extracted; (b) studies conducted on other subtypes of stroke; (c) cohort studies, cross-sectional studies, and randomized controlled trials; and (d) duplicate publications from the same study with overlapping subjects.Extraction of data and evaluation of methodological qualityWe have used the standardized data collection form to extract the data from the included studies. The following important data were extracted for the present study: first author's name, year of article publication, journal in which the article was published, number of genotypes reported in the cases and controls, mean age of cases and controls, and ethnicity. To avoid duplication of the material, we kept only the most recent article or entire study where the same population was reported in multiple publications. Any disputes between the writers were settled through dialogue. For the purposes of the study, ethnicities were divided into two categories: Asian and Caucasian. We also used a quality rating scale created for genetic association studies to assess the methodological quality. Traditional epidemiologic considerations, as well as genetic issues, were included in this scale . The scoPooled odds ratio (OR) with 95% CI was used to determine the pooled effect size . The I2 ResultsPreviously done meta-analysis studies investigating MTHFR C677T polymorphism and ischemic stroke with OR are shown in Table A total of 67 studies that met the inclusion criteria were included in this study, having 17,704 cases and 21,981 controls. The studies were conducted from the period of 1997 to 2020. There were 41 studies from the Asian population and 26 from the Caucasian population. Figure Publication biasThe probabilities of publication bias arising from the published literature were examined using a funnel plot and the Begg’s and Egger's tests. We observed that there was significant publication bias (P < 0.001), indicating that there were probabilities of publication bias Figure .DiscussionOur meta-analysis, which included 67 studies, observed that variation at the C677T position of the MTHFR gene might be associated with an increased risk to develop ischemic stroke.Earlier meta-analyses ,17 with In view of these, our meta-analysis is the largest meta-analysis that used the robust statistical method and methodological quality to derive the precise conclusion regarding the relationship of MTHFR gene polymorphism at 677 positions with the risk of ischemic stroke. In the stratified analysis, the association was found to be higher in the Asian population  as compared to the Caucasian population . However, in meta-regression analysis, ethnicity did not contribute to the significant heterogeneity in the pooled effect size. These findings indicate that similar type of association between MTHFR gene polymorphism and the risk of ischemic stroke in both Asian and Caucasian populations. Our meta-regression analysis to explore the source of variation in effect size did not observe the significant influence of mean age, methodological quality, and year of publication of articles on the pooled effect size. These observations further strengthen the homogeneous effect of the MTHFR gene polymorphism with an increased risk of ischemic stroke.MTHFR polymorphism leads to a higher level of Hcy. Hcy is a sulfur-containing amino acid and its remethylation leads to the formation of methionine. In the remethylation process of methionine, the methyl donor for the conversion of Hcy to methionine is done by the reduction of 5,10-methylenetetrahydrofolate to 5-methyltetrahydrofolate by the enzyme MTHFR. Elevated plasma Hcy levels can occur due to defective remethylation of Hcy to methionine because mutations in the MTHFR gene could lead to decreased activity of the MTHFR enzyme -18. StroThis meta-analysis sustains the notion of the association of MTHFR gene polymorphism with an increased risk of ischemic stroke. The observed pooled effect size had insignificant heterogeneity, which further strengthens the findings observed in the current study. The study is limited by the presence of publication bias. The association of MTHFR gene polymorphism was found to be higher in the Asian population compared to Caucasians. MTHFR gene polymorphism screening may be included in the guidelines for the prevention and screening of subjects with higher susceptibility to stroke.


Previous research suggests a protective effect of vegetable consumption against chronic disease, but the quality of evidence underlying those findings remains uncertain. We applied a Bayesian meta-regression tool to estimate the mean risk function and quantify the quality of evidence for associations between vegetable consumption and ischemic heart disease (IHD), ischemic stroke, hemorrhagic stroke, type 2 diabetes and esophageal cancer. Increasing from no vegetable consumption to the theoretical minimum risk exposure level (306–372 g daily) was associated with a 23.2% decline  in ischemic stroke risk; a 22.9% (13.6–31.3) decline in IHD risk; a 15.9% (1.7–28.1) decline in hemorrhagic stroke risk; a 28.5% (−0.02–51.4) decline in esophageal cancer risk; and a 26.1% (−3.6–48.3) decline in type 2 diabetes risk. We found statistically significant protective effects of vegetable consumption for ischemic stroke (three stars), IHD (two stars), hemorrhagic stroke (two stars) and esophageal cancer (two stars). Including between-study heterogeneity, we did not detect a significant association with type 2 diabetes, corresponding to a one-star rating. Although current evidence supports increased efforts and policies to promote vegetable consumption, remaining uncertainties suggest the need for continued research. A meta-analysis using the Burden of Proof function identified modest evidence supporting a protective role of vegetable consumption against ischemic heart disease, stroke and esophageal cancer but not type 2 diabetes. The 2019 Global Burden of Diseases, Injuries, and Risk Factors Study (GBD) estimated that 429,000  = 340,000–718,000) deaths and 13.0 million (8.25–17.5) disability-adjusted life years (DALYs) were attributable to low vegetable consumption globally in 2019 (ref. 7). The large estimated burden demonstrates the importance of more fully understanding the relationship between vegetable consumption and possible health outcomes and the strength of evidence supporting the understanding of those relationships.Previous research suggests that greater levels of vegetable consumption are associated with lower risk of several adverse health outcomes, including ischemic heart disease (IHD)1; a 16% (95% CI: 10–21) reduction in IHD risk when comparing 0–200 g daily2; and an 8% (95% CI: 4–13) lower incidence of IHD and a 14% (95% CI: 11–17) lower IHD mortality between individuals in the highest versus lowest quantiles of vegetable consumption3. Similar reductions have been estimated for ischemic stroke risk, with Hu et al.4 estimating a relative risk (RR) of 0.86 (95% CI: 0.79–0.93) when comparing each study’s highest and lowest categories of vegetable consumption. Regarding vegetable consumption and type 2 diabetes, previous meta-analyses have offered mixed results. Whereas Schwingshackl et al.5 suggested a statistically significant RR of 0.92 (95% CI: 0.89–0.95) when comparing consumption of 240 g per day versus 0 g per day, Halvorsen et al.8 did not detect a significant association when comparing type 2 diabetes risk between those consuming 200 g per day and 0 g per day 8. Similarly, the results of meta-analyses of vegetable consumption and esophageal squamous cell carcinoma (ESCC) have also been mixed. Whereas Liu et al.6 found significantly lower ESCC risk associated with increased vegetable consumption 6, Vingeliene et al.9 did not detect a significant relationship 9. Other meta-analyses have found significant associations between vegetable consumption and esophageal adenocarcinoma risk, with an RR of 0.89 (95% CI: 0.80–0.99) for 100 g per day of vegetable consumption versus 0 g per day9 and an RR of 0.76 (95% CI: 0.59–0.96) for the highest versus lowest vegetable consumption groups10.Prior meta-analyses have estimated an 18% : 8–17) reduction in IHD risk when comparing 0–400 g of vegetables daily11 addressed the question of evidence quality and applied the Grading of Recommendations Assessment, Development and Evaluation (GRADE) framework to provide a qualitative summary rating of the evidence. They rated the quality of the evidence for the relationship between vegetable consumption and IHD mortality risk as ‘moderate’ and the evidence for the relationship between vegetable consumption and IHD incidence risk as ‘low’3. The GRADE framework takes into account bias in the data, imprecision in estimation for a particular study, inconsistency among studies, indirectness of the research question of the included study to the research question of study at hand and publication bias11. However, the framework is both qualitative and subjective and, thus, has limited utility for comparing the strength of evidence between risk factors being evaluated by different groups of researchers. Most recently, Schwingshackl et al.12 developed NutriGrade, a nutrition-specific scoring system to evaluate quality of evidence tailored to deal with the particular challenges of nutrition studies that includes assessment of risk of bias, study quality, study limitations, precision, heterogeneity, directness, publication bias, funding bias, study design, effect size and dose response. NutriGrade is a promising approach but uses a scoring system that assigns points with limited empirical basis, and, with its orientation toward nutrition research, it may not be readily applicable or appropriate for evaluating the larger collection of risk factors included in the GBD study. The Hierarchies of Evidence Applied to Lifestyle Medicine (HEALM) was developed as a construct to evaluate the strength of evidence for social, behavioral and genetic risk factors for which randomized trials are impossible, impractical or unethical13. Although offering a useful construct for considering evidence beyond trials, HEALM relies largely on subjective expert assessments and is not applicable or appropriate for all risk factors included in the GBD study. The Newcastle-Ottawa scale10 provides a framework for assessing the risk of bias of observational studies with regard to participant selection and comparability and the ascertainment of exposures and outcomes. This framework is useful, but, similarly to the GRADE framework, it faces limitations as a subjective and qualitative assessment with poor inter-rater reliability and does not offer a mechanism to synthesize the quality and strength of evidence across all relevant studies to generate a global assessment14.An understanding of the direction and magnitude of associations between vegetable consumption and adverse health outcomes, as well as the strength of the evidence supporting the existence of these associations, is essential for guiding policy and individual health decisions related to diet and agriculture. Although many of the meta-analyses provide 95% CIs on their dose–response curves, this statistical uncertainty does not capture uncertainty that might stem from differences in study design , potential publication bias, between-study heterogeneity or small numbers of studies. Zurbau et al.The standard meta-analytic approach of reporting mean effects offers an objective and quantitative measure of the strength of associations observed in epidemiological studies, but it offers no assessment of the extent to which we are to believe that the mean effect accurately represents the true association versus a manifestation of biases and errors. Conversely, existing methods to judge the quality of the evidence  offer a means to assess the quality of the underlying evidence, but they do so subjectively and do not extend to estimating what the true effect size might be in light of any biases or other limitations of the evidence.15 developed a Bayesian meta-regression tool and methods for assessing the quality of evidence for risk–outcome associations that are both objective and quantitative15. This approach is intended to complement and bridge the gap between quantitative meta-analyses and subjective assessments of evidentiary strength. It offers a means to evaluate the strength of epidemiological evidence for associations between risk–outcome pairs using an objective and empirical approach within a transparent regression framework and connects this to the estimate of the mean effect by reporting the strength of the association that is consistent with the most conservative interpretation of the evidence. A burden of proof risk function (BPRF) can then be calculated from this conservative risk function as the 5th (for harmful) or 95th quantile risk curve that is closest to the null, and this BRPF represents the weakest association across the range of exposures that is consistent with the available data: the farther the BRPF is from the null, the stronger the evidence is for that association16. Finally, the exposure-averaged BPRF for a risk–outcome pair is converted into a risk–outcome score (ROS) that can be compared across risk–outcome pairs and assigned a star rating (from one to five) based on the quantitative assessment of the association15.As part of the GBD study, Zheng et al.17 and for which there existed sufficient data to enable estimation of all components of the modeling chain. The main findings and policy implications of the current study are summarized in Table For this study, we applied this meta-regression approach to both quantify health effects of vegetable consumption and evaluate the quality of the evidence supporting those effects. We conducted a systematic review to identify studies that estimated associations between vegetable consumption and the five health outcomes for which we estimate burden attributable to low vegetable consumption in the GBD study: type 2 diabetes, esophageal cancer, hemorrhagic stroke, IHD and ischemic stroke. These outcomes were selected across previous iterations of the GBD study as those for which there existed convincing or probable evidence for a causal effect based on the World Cancer Research Fund Grading System18, we systematically searched the literature for studies reporting associations between vegetable consumption and each of the five included health outcomes, and we screened records based on pre-defined inclusion and exclusion criteria . Data collection across studies occurred between 1967 and 2020 and included 4,609,296 participants from 34 countries and territories  guidelinescriteria . The revcriteria . We idenIncreased vegetable consumption was associated with reduced risk of all included outcomes, and we subsequently selected the theoretical minimum risk exposure level (TMREL) to correspond to high real-world consumption levels for which we had adequate data to draw robust conclusions about risk. This TMREL was based on observed exposure levels reported in the included studies . Six cohorts were from Europe27; four cohorts were from Asia68; and two cohorts were from North America29. Ten studies determined outcomes using administrative medical records or disease registries29; one study used self-reported incidence19; and one study used death certificates68. In all studies, the effect size measure was adjusted for body mass index (BMI), whereas eight studies adjusted for type 2 diabetes68; 11 studies adjusted for physical activity29; six studies adjusted for hypertension68; and ten studies adjusted for alcohol intake68. Seven studies reported effect sizes with hazard ratios (HRs)68, and the rest reported RRs28.We identified 12 prospective cohort studiesWe observed a statistically significant, non-linear, monotonic decline in ischemic stroke risk associated with higher vegetable consumption Fig. . The larA conservative (closest to null) interpretation of the evidence suggests that the average vegetable intake  was associated with a 14.2% lower risk of ischemic stroke compared to 0 g of vegetable intake. We converted this BPRF to an ROS of 0.15, corresponding to a three-star rating. Ratings of two through five translate to associations that remain statistically significant after accounting for between-study heterogeneity, with a three-star rating indicating that average exposure decreases risk by 13–34% compared to no exposure. Our analysis did not find significant evidence of publication bias, and visual inspection of the funnel plot provides no evidence of substantial publication bias Fig. .67, including a total of 1,952,226 participants and 41,494 IHD events. Five studies used IHD incidence61 as the endpoint to estimate effect sizes; four studies used mortality57; and eight studies used both IHD incidence and mortality67. Fourteen studies determined outcomes using administrative medical records, disease registries or death certificates67; one study supplemented records with self-reported incidence42; and two studies used physician diagnosis60. All studies but one adjusted their effect size measure for age67, and ten studies either adjusted for or stratified by sex67. Other common adjustment variables include smoking (n = 16)69, alcohol consumption (n = 14)67, BMI (n = 15)67, physical activity (n = 15)67, education level (n = 11)67, hypertension (n = 9)60 and type 2 diabetes (n = 8)60. Twelve studies reported effect sizes with HRs67, and five reported RRs50.Our analysis of vegetable consumption and IHD included 78 observations from 17 prospective cohort studiesWe observed a statistically significant, non-linear monotonic decline in IHD risk associated with higher vegetable consumption Fig. . The larCompared to the TMREL (306–372 g per day), consuming no vegetables was associated with a 22.9%  greater mean risk of IHD. Notably, the uncertainty around the risk curve is modest, and we observed little difference in uncertainty whether the effect of between-study heterogeneity was included or not .Using the BPRF analysis, we found that average vegetable consumption levels  were associated with a 12.1% decreased risk of IHD compared to 0 g of consumption. This is a conservative estimate that corresponds to the smallest reduction in risk (closest to null) that is consistent with the available evidence and inclusive of between-study heterogeneity and other forms of uncertainty. This BPRF equated to an ROS of 0.13, corresponding to a two-star rating. Ratings of two through five translate to associations that remain statistically significant after accounting for between-study heterogeneity, with a two-star rating indicating that average exposure decreases risk by 0–13% compared to no exposure. We found that trimming had little effect on the ROS, and, without trimming, the score increased to 0.127, suggesting that the ROS for vegetable consumption and IHD is relatively insensitive to outliers. Although we observed little asymmetry in the funnel plot Fig. , the res68, including a total of 646,338 participants. Eight studies determined outcomes using administrative medical records68, disease registries or death certificates, and one study used self-reported incidence19. All studies adjusted for age; six studies adjusted for other components of diet68; and eight studies either adjusted for or stratified by sex68. Other common adjustment variables included smoking (n = 9)69, alcohol consumption (n = 7)68, BMI (n = 9)69, physical activity (n = 8)27, education level (n = 4)27, hypertension (n = 5)68 and type 2 diabetes (n = 7)68.Our analysis of vegetable consumption and hemorrhagic stroke included 30 observations from nine prospective cohort studiesWe observed a statistically significant, non-linear, monotonic decline in hemorrhagic stroke risk with increasing vegetable consumption Fig. . The larCompared to the TMREL, consuming no vegetables was associated with a 15.9% (1.7–28.1) greater risk of hemorrhagic stroke. The ROS was 0.016, indicating a two-star rating. We did not find significant evidence of publication bias, and visual inspection of the funnel plot provides no evidence of substantial publication bias Fig. .62 to evaluate the relationship between vegetable consumption and esophageal cancer, with a total sample size of 818,059 individuals. The follow-up duration ranged from 5 years to 16.3 years, with a median duration of 8.9 years. Three of the cohorts were from Asia62; two cohorts were from Europe56; and one cohort was from the United States35. Four of the cohorts used administrative medical records to assess the outcome59, whereas the other two cohorts used physician diagnosis62 or self-report55. Four of the cohorts used HRs59 as the effect size measure, and two of the cohorts used RRs66. All but one study adjusted the RR for smoking59.We identified six prospective cohort studiesWe observed a statistically significant, non-linear monotonic decline in esophageal cancer risk with increasing vegetable consumption Fig. . UniquelP = 0.031).Compared to the TMREL, consuming no vegetables was associated with a 28.5% (−0.02 to 51.4) greater risk of esophageal cancer. The ROS was 0.001, corresponding to a two-star rating. We observed asymmetry in the funnel plot Fig. , and the65 and one case–cohort study63 evaluating the relationship between vegetable consumption and type 2 diabetes among 1,025,899 participants. Six cohorts64 were from Europe; five cohorts were from North America51; four cohorts were from Asia65; and two cohorts were from Australia52. Ten studies used self-report65 to assess outcomes; three studies used administrative medical records63; two studies used physician diagnosis33; and one study used biomarkers to assess the outcome of interest30. Follow-up time ranged from 4 years to 23 years (median: 10.6 years). In all studies, the effect size measure was adjusted for BMI, 12 studies adjusted for education65, and six studies adjusted for hypertension65. Eight studies reported effect sizes with HRs65; four studies reported ORs64; and the rest reported RR.We identified 15 prospective cohort studiesWe observed a non-linear, monotonic decline in type 2 diabetes risk with increasing vegetable consumption that did not achieve statistical significance when accounting for between-study heterogeneity Fig. . Compare15 both to quantify the association between vegetable consumption and five health outcomes and to assess the strength of the evidence for those associations. We found that increased vegetable consumption was associated with statistically significant reductions in the risk of ischemic stroke risk, and that, although evidence is suggestive of protective effects with type 2 diabetes, the evidence did not achieve statistical significance when between-study heterogeneity was accounted for. For all outcomes except esophageal cancer, we estimated that mean risk decreased non-linearly with increased consumption such that the most substantial declines occurred with increases from low to moderate consumption, and smaller marginal risk reduction occurred with increases from moderate to high consumption.We applied our methodological framework15 . These findings suggest that we have stronger evidence to support the associations between vegetable consumption and both IHD and ischemic stroke than we do for most risk–outcome pairs in GBD but that the evidence is not overwhelming. Although the BPRF star and GRADE ratings are not directly comparable, it is notable that both approaches suggest that evidence with regard to IHD is moderately good but not of the highest quality3. Conversely, the two-star rating for IHD suggests weaker evidence than does the World Health Organization (WHO)/Food and Agriculture Organization (FAO) assessment of ‘convincing’ evidence for decreased risk of cardiovascular disease associated with fruits and vegetables, which corresponds to ‘evidence based on epidemiological studies showing consistent associations’70. This difference highlights the importance of considering the strength of the association and using a quantitative approach to assessing consistency  when evaluating evidence.We found ROSs of 0.15 and 0.13 for ischemic stroke and IHD, respectively, corresponding to a three-star rating and a two-star rating for two outcomes , respectively. These ratings reflect the reasonably good quality and number of studies, large sample sizes and little between-study heterogeneity on the one hand and modest magnitude of the effect sizes on the other: all else being equal, risk–outcome pairs with weaker associations will have lower ROSs to reflect the greater potential for weak associations to result entirely from residual confounding or bias. For context, of the 180 risk–outcome pairs assessed with this method to date, 46 (25.6%) received a three-star rating; 112 (62.2%) received a rating of one or two stars; and only 22 (12.2%) received a rating of four or five stars71. Evidence for an association between vegetable consumption and type 2 diabetes is still weaker, with a one-star rating. Although we found evidence trending toward a protective effect of some vegetable intake compared to none, the association did not achieve statistical significance using an approach to capturing uncertainty that accounts for between-study heterogeneity. For type 2 diabetes, we found statistically significant associations with vegetable consumption when not accounting for between-study heterogeneity, indicating that this lack of consistent findings across studies is a major factor underlying the weak ROS. The WHO and FAO did not include an assessment of the evidence for an association between vegetables in totality and type 2 diabetes; they assessed a protective effect of fruits and vegetables against type 2 diabetes as ‘probable’ based on evidence for non-starch polysaccharides, for which fruits and vegetables are a major source. The lack of clarity around vegetables and type 2 diabetes risk suggests a need for more research.Evidence for associations between vegetable consumption and hemorrhagic stroke and esophageal cancer is weaker, with each having a two-star rating. We found statistically significant evidence both with and without between-study heterogeneity for each of these two outcomes, but the strength of that evidence is weakened by the modest effect sizes, relatively limited number of studies and large between-study heterogeneity. The two-star rating for esophageal cancer is in line with, although perhaps somewhat stronger than, the assessment of the evidence as ‘limited, suggestive decreases risk’ by the World Cancer Research Fund and the American Institute for Cancer Research1 and the present study are 0.82 (95% CI: 0.73–0.92) and 0.76 (0.68–0.86), respectively; and comparing 200 g to no vegetable consumption, the RR estimates from Aune et al.2 and the present study are 0.84 (95% CI: 0.79–0.90) and 0.80 (0.72–0.88), respectively. Similarly, we estimated a stronger effect size with type 2 diabetes than did previous meta-analyses. Whereas Schwingshackl et al.5 estimated an RR of 0.92 (95% CI: 0.89–0.95) comparing consumption of 240 g per day versus 0 g per day and Halvorsen et al.8 estimated an RR 0.97 (95% CI: 0.94–1.01) comparing 200 g per day and 0 g per day, we estimated an RR of 0.73 (0.52–1.04). Note that, although our estimate suggests a stronger mean effect size, we also estimate much greater uncertainty, and our UIs include both previously published estimates. Previous meta-analyses of vegetable consumption and esophageal cancer have been mixed, and most estimated RRs by esophageal cancer type , preventing direct comparison. Vingeliene et al.9 looked at vegetables and all esophageal cancer and estimated an RR of 0.98 (95% CI: 0.90–1.06) comparing 100 g to 0 g of vegetables daily, whereas we estimated a corresponding RR of 0.91 (0.84–1.00). As with the aforementioned outcomes, our estimates suggest a slightly, but not significantly, stronger effect size.Our analysis suggests slightly, but not significantly, stronger protective effects of vegetables against IHD than reported in previous meta-analyses. For example, comparing 400 g to no vegetable consumption, the RR estimates from Gan et al.72, we expect strong confounding with any naturally observed association between vegetables and health in the absence of thoughtful and well-executed statistical controls. Broadly, we expect vegetable consumption to be positively associated with more healthful behaviors  and with demographic and social factors associated with improved health . This suggests that confounding would distort the strength of the association upward . Although all of the studies in this analysis adjusted for some potential confounders, it is likely that some residual confounding remains due to imperfect statistical adjustment ; imperfect confounder measurement; and the existence of confounders that were unknown, unmeasured or otherwise unadjusted for. This is one reason that the BPRF methodology rates evidence of similar quality more strongly for risk–outcome pairs for which we observed a stronger association: strong associations are less likely to result entirely from residual confounding (and other sources of bias) than are weak associations16. Although we included bias covariates for imprecise exposure definitions, there are some potential sources of exposure misclassification for which we were unable to completely account: errors in measuring dietary intake are well-established due to self-reporting errors, random daily variation in diet and differences in perceptions of portion sizes69. Many studies are careful to define a serving size when collecting and reporting data; where studies did not define a serving size, however, we applied a consistent definition that could differ from perceptions or definitions of a serving size within the study context. For example, a dietary assessment calibration study in ten European countries found that participant definitions of a serving of vegetables ranged from 43.8 g in Sweden to 89.6 g in Greece73. Although random errors in exposure measurement will bias RR estimates toward the null, the effect of systematic errors are harder to predict. Finally, this study is subject to the limitations of the methodological framework15.Although the methodological framework used in our work addresses many of the limitations of existing approaches, our study nonetheless has several important limitations. We identified no randomized trials of the effect of increased vegetable consumption on any of the five health outcomes. Because vegetable consumption is strongly associated with other aspects of diet and other demographic, social and behavioral risk factors The Lancet EAT Commission recommends that people consume a primarily plant-based diet, including 300 g of vegetables per day to promote both individual human health and environmental health74. Although our analyses did not evaluate the potential environmental benefits of increased vegetable consumption, our results do support the recommendation for increased vegetable consumption as it relates to individual-level health benefits, and our TMREL estimates of 306–372 g per day are in line with the Commission’s recommendation. Several other dietary guidelines express recommendations for fruits and vegetables combined: both the WHO and the World Cancer Research Fund recommend consuming at least 400 g of fruits and vegetables daily75, and the National Health Service of England’s Eat Well guide recommends at least 385 g of combined fruits and vegetables76. The United States Department of Agriculture’s Dietary Guidelines for Americans recommends 2.5 cups of vegetables daily, but both the volume-based measure and the inclusion of beans, lentils and starchy vegetables makes direct comparison to our estimates difficult77.Based on our analysis, the evidence justifies more robust efforts and policies to promote increased vegetable consumption to reduce chronic disease risk, especially with regard to IHD, stroke and esophageal cancer. At the population level, increased vegetable consumption is also one component of a diet that is favorable for sustained planetary health. Given the large carbon footprint associated with the production of animal-based foods, We found evidence that increasing vegetable consumption is associated with a reduction in IHD, ischemic stroke, hemorrhagic stroke and esophageal cancer risk, and we found a TMREL that is in line with existing recommendations for vegetable consumption. The ROSs for ischemic stroke and IHD, in particular, are among the highest for the dietary risk factors that have been evaluated to date, and the associated two-star and three-star ratings indicate that, even under a conservative interpretation of the evidence, vegetable consumption has a significant protective effect against ischemic stroke, IHD, hemorrhagic stroke and esophageal cancer risk. Although randomized controlled trials (RCTs) would help to definitively clarify the extent to which residual confounding and bias may distort estimates from observational studies, current evidence justifies more robust efforts and policies to promote increased vegetable consumption.16, we applied the meta-analytic approach detailed in Zheng et al.15 to evaluate the magnitude and strength of evidence for the association between vegetable intake and five potential health outcomes. This approach can be summarized in six steps: (1) search for and extract data from published studies; (2) estimate the shape of the exposure-relative risk relationship; (3) test and adjust for biases across study designs and characteristics; (4) quantify unexplained between-study heterogeneity; (5) evaluate potential small-study effects and publication or reporting bias; and (6) estimate the BPRF—a quantification of the smallest harmful or protective effect (closest to null) across a range of exposures that is supported by the evidence—and use this estimate to compute the ROS and star rating on a scale of one to five. We describe these steps below, and further details are available in Zheng et al.15.Using tools first published in Zheng et al.The estimates for our primary indicators from this work—RRs across a range of exposures, BPRFs, ROSs and star ratings for each risk–outcome pair—are not specific to or disaggregated by specific populations . The reports we referenced included information about the self-reported sex of the participants but did not all include sex-specific RR estimates. These factors precluded us from performing any sex-based or gender-based analyses.18  between vegetable consumption and each of the five included health outcomes. We define vegetable consumption as the average daily consumption (in grams per day) of vegetables, including fresh, frozen, cooked, canned or dried vegetables, and excluding legumes and salted or pickled vegetables, juices, nuts and seeds and starchy vegetables, such as potatoes or corn. We defined our outcome as either incidence of, or mortality from, the health outcome, excluding studies that included other or non-specific outcome definitions .79, stroke subtypes79 and type 2 diabetes8. Search strings were developed to identify the meta-analyses for each outcome. We searched the citations of each identified meta-analysis. A separate search string was developed to identify sources published after the period covered in the identified most recent PRISMA-compliant meta-analysis for each outcome of interest. We searched PubMed, EMBASE and Web of Science from the last date of the identified meta-analysis through 31 May 2022. We also searched the Global Health Data Exchange (GHDx) databases. For vegetable consumption and esophageal cancer, we conducted a full systematic review with no date constraint because we did not find a meta-analysis that fulfilled our selection criteria (PRISMA-complaint and matching our definition of exposure and outcome). A detailed description of the search strings and searching strategy is presented in the Supplementary Appendix. After removing duplicate search results, reviewers manually screened each study for eligibility on the basis of title and abstract. As a sensitivity check, 20% of title and abstract exclusions by each reviewer were verified by a second reviewer. The full text of the remaining studies was retrieved and examined for inclusion in this analysis. Finally, each full-text exclusion was verified by both reviewers. Only one discrepancy was found and was settled by moving to full-text review for additional evaluation.We first searched the most recent PRISMA-compliant meta-analyses that examined associations between vegetable consumption and IHD80. For studies that reported mean consumption rather than ranges of vegetable consumption, we used the midpoint between means as the cutoff for intake intervals. For undefined lower bounds, we assumed a consumption level of 0 g per day. For undefined upper bounds when the mean and standard deviation values were not available, we applied the range from the cohort’s most adjacent quartile or tertile to estimate the upper bound of consumption, specific to each study cohort. All studies included in the analysis for each risk–outcome pair and study characteristics are given in Supplementary Table Studies were eligible for inclusion if they used suitable exposure and outcome definitions, provided some measure of uncertainty  and reported an RR (or related measure) for which the exposed and unexposed groups were defined. Studies that estimated associations with a select subset of vegetables  were excluded. Where multiple studies provided RR estimates from the same cohort, we included only the study that captured the largest sample or follow-up time so as not to include duplicate data. For each study, two reviewers manually extracted data on study name, location, design, population , duration of follow-up, exposure definition, exposure assessment method, exposure categories, outcome definition, outcome ascertainment method and covariates included in the statistical analysis of the study. All included studies published the required data, and no unpublished data were obtained for this analysis. For each exposure category, we also collected data on the range of exposure, number of participants, person-years, number of events and risk estimate and its corresponding uncertainty. See Supplementary Table 16. MR-BRT has capabilities that are useful for estimating risk curves and assessing the quality of evidence for the modeled association, including (1) the ability to model non-linear associations with splines with automatic knot selection and shape constraints; (2) the ability to use study-level covariates to account for elements of study design that could introduce bias; (3) the ability to estimate uncertainty with and without consideration of between-study heterogeneity; (4) the ability to adjust the between-study heterogeneity parameter to penalize data sparsity through a Fisher scoring correction; and (5) the ability to assess publication bias15.We modeled a mean RR (a measure of effect size) curve across the range of exposure for each risk–outcome pair using MR-BRT, a Bayesian meta-regression tool detailed previously81; (2) a one-stage approach for dose–response meta-analysis82, implemented in the dosremeta package83; and (3) a two-stage approach for dose–response meta-analysis84, implemented in the dosremeta package83. The approach used here performed as well as or better than existing approaches across all three scenarios, particularly for non-linear relationships15.We followed a consistent approach to choosing model specifications for all dietary risks. We first modeled the association with no constraints to assess the nature of the association and classified risk curves as protective (risk declining with increasing exposure), harmful (risk increasing with increasing exposure) or J-shaped . We then applied the same number of spline degrees and knots and same priors and constraints to all risk–outcome pairs with a given shape. In the case of vegetable consumption and each of the five outcomes presented here, our unconstrained models suggested that disease risk declined with increasing vegetable consumption, corresponding to a protective risk. As a protective risk, we ran the final models using quadratic splines with two internal knots, a monotonic decreasing prior and a linearity prior on the right tail. Major components of the meta-regression model were validated before their use in this study, including simulating scenarios for (1) many studies with many data points per study; (2) many studies with few data points per study; and (3) few studies with few data points per study. For each simulation, results from the approach used here were compared with results from three existing approaches: (1) a log-linear meta-analysis implemented in the meta-for package15. See Supplementary Table For each study reporting an effect size for the association between vegetable consumption and the included outcomes, we extracted information about aspects of study design that could potentially bias the reported effect size and coded this information into study-level covariates. These study-level covariates included follow-up period (≤10 months and >10 months), precision of the exposure and outcome definitions, study design , reported measure of association (RRs or ORs), included outcome events , number of exposure measurements (single or repeat), method by which outcomes were ascertained (administrative records or self-reports) and the level of adjustment for relevant confounders, such as age, sex, smoking, education, calorie intake and income. We adjusted for these covariates in our meta-regression if they significantly biased our estimated RR function15. As described in Zheng et al.15, we implemented the Fisher information matrix to the heterogeneity parameter, which corrects potential underestimation of between-study heterogeneity for data-sparse situations. In such cases, the between-study heterogeneity parameter estimate may be zero, simply from lack of data. The Fisher information matrix uses a quantile of gamma, which is sensitive to the number of studies, study design and reported uncertainty. We calculated 95% UIs for each mean risk curve both with between-study heterogeneity incorporated (a ‘conservative’ UI) and without it incorporated , but we present only the UIs that include between-study heterogeneity.Heterogeneity in RR estimates among different studies may reflect methodological differences, random variation or true inconsistencies in effect sizes in different study populations. We used the aforementioned study-level covariates to account for between-study heterogeneity that results from differences in study design and used a linear mixed-effects model to capture the remaining unexplained between-study heterogeneityP value to assess the statistical significance of that association, where the residual standard deviation incorporates both the standard error of the data point and between-study heterogeneity85. Small-study biases may arise from random change , differences in study quality or analysis or publication and reporting biases. The formal statistical test used here decreases the chance of flagging apparent small-study biases that arise only from random chance; and our application of bias covariates in the RR model decreases the change of systematic differences in model residuals arising from differences in study quality or analysis. In this context, Egger’s regression becomes a tool primarily to test for empirical evidence of potential publication or reporting biases85 that identifies significant evidence of publication bias as a statistically significant association between model residuals and the residual standard deviations of the corresponding data points. Significant results for Egger’s regression are flagged as indicating a potential risk for publication and/or reporting bias15.Egger’s regression assesses small-study bias by estimating the association between model residuals and the residual standard deviation of the corresponding data points, and the The TMREL is the level of exposure that, within the range of values that are theoretically possible at the population level, will minimize the risk of all outcomes associated with that risk combined. For harmful exposures that may be theoretically eliminated, the TMREL is typically zero ; and for exposures with U-shaped risk curves, the TMREL is typically set to the exposure level that corresponds to the minimum of the risk curve . Defining the TMREL for protective exposures, however, is more conceptually difficult and requires us to consider three factors: 1) we cannot assume that marginal protective effects persist with increases in exposure beyond the exposure levels that we observe in the source data ; 2) for most protective exposures, the risk curve will be steepest at lower levels of exposure and flatten at higher exposure levels, leading to risk reductions at higher levels that are neither statistically nor clinically significant; and 3) as we move toward very high exposure levels, the requirement for the TMREL to be theoretically possible becomes relevant and supports basing the TMREL on plausibility. We, therefore, selected the TMREL based on exposure levels reported in the included RR studies. For each study, we took the range of vegetable consumption for the highest consumption group. We defined the lower bound of the TMREL as the 85th percentile of the lower limit of that range across all studies, and we defined the upper bound of the TMREL as the 85th percentile of the midpoint of the range across all studies. This approach yields a TMREL that corresponds to high real-world consumption levels for which we have adequate data to draw robust conclusions about risk.16. We then estimated the ROS as the mean value of the log-BPRF averaged over the 15th and 85th percentiles of the distribution of vegetable consumption observed in the studies used to estimate the association between vegetable consumption and the outcome of interest. Because log RRs are symmetric about zero, with harmful risks having positive values for the log RR and protective effects having negative values, we multiplied the ROS by –1 for each risk–outcome pair in this analysis, because vegetable intake is a protective risk. The ROS, therefore, provides a single summary of the log RR that is comparable across both protective and harmful effects. A higher positive ROS corresponds to a stronger association, whereas a negative ROS indicates that the available evidence did not reject the null hypothesis. Finally, we converted the ROS for each risk–outcome pair into a star rating on a scale of one to five, with a one-star rating indicating a non-significant relationship based on the conservative interpretation and two-star through five-star ratings representing an increase in risk with average exposure (compared to no exposure) in the ranges of 0–15% for two-star pairs, >15–50% for three-star pairs, >50–85% for four-star pairs and greater than 85% for five-star pairs for harmful risks and a decrease in risk with average exposure (compared to no exposure) in the ranges of 0–13% for two-star pairs, >13–34% for three-star pairs, >34–46% for four-star pairs and greater than 46% for five-star pairs for protective risks15.Using the risk curve modeled with the MR-BRT tool and incorporating between-study heterogeneity into our estimate of uncertainty, we estimated the BPRF as the function that corresponds to the 5th (for harmful risks) or 95th (for protective risks) quantile of the log RR curve that is closest to the null. The BPRF represents the weakest association that is consistent with the available data after accounting for between-study heterogeneity, and the further the BPRF falls from the null, the stronger the evidence is for that association15). We also conducted additional sensitivity analysis by including calorie intake adjustment as a pre-selected bias covariate. The dose–response curves of the sensitivity analyses are shown in Supplementary Figs. For each risk–outcome pair, we conducted sensitivity analyses that compared the dose–response curves generated with and without trimming the 10% least coherent data points (online methods section 8 in Zheng et al.Analyses were performed using R version 3.6.1, Python version 3.8 and Stata version 17.This study was a secondary analysis of existing data involving systematic reviews and meta-analyses. No statistical method was used to pre-determine sample size. As the study did not involve primary data collection, randomization, blinding and data exclusions are not relevant to this study and, as such, no data were excluded, and we performed no randomization or blinding. We have made our data and code available to foster reproducibility.Further information on research design is available in the Any methods, additional references, Nature Research reporting summaries, source data, extended data, supplementary information, acknowledgements, peer review information; details of author contributions and competing interests; and statements of data and code availability are available at 10.1038/s41591-022-01970-5.Supplementary InformationReporting Summary


Background and Objectives: The aim of this study was to investigate the association between smoking status and single-nucleotide polymorphism in candidate genes that had a known association with smoking-related tumors in previous studies and to explore their link to laryngeal cancer risk in a population of northern Romanian descent. The genes selected have key functions in xenobiotic metabolism (GSTs: the glutathione S-transferases family: GSTM1 and GSTT1) and chromosomal management (TERT). Materials and Methods: The genotype frequencies of TERTRs2736100 and the GST subfamilies (GSTM1 and GSTT1) were determined using the polymerase chain reaction-restriction fragment length polymorphism (PCR-RFLP) technique. The relationship between the polymorphisms and the risk of laryngeal cancer was analyzed in a retrospective case–control study of 92 laryngeal cancer cases and 101 controls, all of whom were smokers. Results: Subjects presenting the GSTT1-null variant had a two-fold increase in risk . While no individual risk was observed for the TERTRs2736100 polymorphism, stratification based on gender revealed a nine-fold increase in risk for carriers of the “C” allele in the heterozygote variant who were male . Conclusions: The results showed that the GSTT1-null genotype and the mutant heterozygote variant of TERTRs2736100 genes may play a significant role in laryngeal cancer susceptibility in subjects of northern Romanian descent. There may be no association between the susceptibility to laryngeal carcinoma and the GSTM1 polymorphism. The results could not confirm the carcinogenic influence smoking has on laryngeal cancer development for the studied polymorphisms. To date, carcinogenesis has been linked with genetic and environmental/lifestyle factors. However, the multitude of possible combinations between them has made the identification of the exact etiology a continuous struggle ,2. ScholEpidemiological studies focusing on the development of this malignancy have identified a wide range of influential factors ,7, amongConcurrently, the aforementioned risk tends to vary greatly depending on the number of cigarettes consumed, the duration of smoking, and the tumor subsites ,10,11. IThe characterization of laryngeal cancer through genetic testing over the years has led to the identification of several candidate genes with key functions linked to the carcinogenic process in populations around the world . HoweverIn this study, we performed a retrospective case–control study on 92 LSCC subjects and 101 controls, all smokers, from the Department of Genetics and the Department of Molecular Sciences, “Iuliu Haţieganu” University of Medicine and Pharmacy, Cluj-Napoca, Cluj County, Romania. Details regarding the age, gender, and smoking duration (in years) of the subjects were recorded. Grouping according to yearly consumption was carried out according to the mean value: light smokers (<29 years) and heavy smokers (≥29 years).The polymerase chain reaction-restriction fragment length polymorphism (PCR-RFLP) technique was used to identify the genotype frequencies of the three candidate polymorphisms. DNA was extracted from ethylenediaminetetraacetic acid (EDTA) blood using specific DNA extraction kits and put through PCR fragment amplification and successive thermocycling conditions  according to the manufacturer’s protocol .p) did not exceed 0.05. OR and CI were not calculated when data were too sparse. The final results were then subjected to false discovery rate verification.The chi-square test and Fisher’s exact test were performed to analyze the established categorical data sets . Logistic regression models were used to identify possible associations between individual genotypes and outcome parameters. The analysis was initially carried out using the polymorphism of each gene as a predictor variable and subsequently in combination with other predictor variables. All genotypes were converted into dummy variables for this process. Statistical analysis was carried out using the Epiinfo (version 7.2.3.1) software. Crude and adjusted odds ratios (OR) with corresponding 95% confidence intervals (CIs) were calculated to estimate the risk and strength of association. The homozygote wild-type genotype of each polymorphism was used as the reference category for comparative analysis. Results were considered significant if the value of probability (p = 0.41). The predominant age group consisted of subjects over 60 years of age without a significant difference between cases and controls (p = 0.06). Among the smoking population, the most common group consisted of moderate smokers, but the difference was not statistically significant (p = 0.14).The aforementioned study population was divided according to age, gender, and tobacco consumption, as reported in Rs2736100, GSTM1, and GSTT1 are reported in Rs2736100, the mutant homozygote variant CC was the most common (39.13%) in cases, while the mutant heterozygote variant AC appeared more frequently in controls (42.57%). Though the GSTM1-null genotype was more common in laryngeal cancer subjects than in controls (55. 43% vs. 48.51%), as opposed to the GSTM1-positive genotype, which showed a reverse distribution (51.49% vs. 44.57%), the results were not statistically significant (p = 0.38). The GSTT1-positive genotype was more frequent in controls (80.20%) than in cases (66.30%), and the difference was statistically significant (p = 0.03). As to the concurrent GSTM1/GSTT1 genotypes, the dual positive genotype had the highest frequency among controls (42.57%), while the GSTM1-null/GSTT1-positive genotype had the second-highest distribution among both cases and controls. However, no statistically significant difference in the genotype and allele distribution was identified.The identified genotypes and allele frequencies of TERTRs2736100, GSTM1, and concurrent GSTM1/GSTT1 genotypes in any of the studied genetic models. However, GSTT1-null showed a two-fold increase as a protective cancer factor when compared to the reference genotype .The comparative analysis of genetic models for the variant carriers of laryngeal cancer risk is shown in p = 0.00000) compared to the reference category  as potential risk factors for laryngeal cancer among smokers of northern Romanian descent. Studies on the existing relationships between genetic polymorphisms and cancer development have been extensive, although the findings are yet to be fully conclusive. The applicability of TERTRs2736100 has been well documented in connection to the neoplastic process [This study is the first to investigate the polymorphisms of TERT process ,16 due t process ,18. Unde process . Meta-an process ,21, whil process  and myel process .Rs2736100 and laryngeal cancer as an independent risk factor for any of the individual genotypes. However, subsequent stratification based on gender highlighted the association between laryngeal cancer and the mutant heterozygote genotype AC. Males carrying at least the “C” allele were at far higher risk for LSCC than those without . This result confirms the importance of gender distribution in neoplastic development [Rs2736100 and laryngeal cancer susceptibility, contrary to those studies reporting that a minimum period of smoking ranging from 5 to 30 years caused a high neoplastic risk and that tobacco smoke was a stable and independent major risk factor [Among the studied population, the results of our research could not find any positive association between TERTelopment  performek factor ,12,24,25p = 0.04), and Lourenço et al. [Due to their fundamental role in the detoxification process of carcinogenic reactive metabolites, the literature has characterized GSTM1 (class Mu) and GSTT1 (class Theta)—of the GST supergene family—as carcinogen-metabolizing genes . In heado et al.  noted thThe results of the present study could not corroborate this particularity of the GSTM1 gene for our Romanian population group. However, our analysis of the GSTT1-null genotype highlights the implications that deletion has concerning its influence on cancer progression within this pathology. When applying the same stratification based on smoking status, we could not validate the notion that tobacco consumption appears to boost individual sensitivity as patients with reduced detoxification cannot operate on the same level in terms of expelling the toxic compounds which characterize tobacco intake . In oppoBecause this is the first study to include the examination of these specific genes in a Romanian population group, we consider our results as preliminary to further independent studies required to verify our results with a larger sample size/geographic distribution within the territory of Romania.Our results highlight the importance of deletion in GSTT1 as an independent risk factor for neoplastic development in laryngeal cancer. Carriers of the TERT Rs2736100 mutant heterozygote genotype (AC) have a significantly higher risk of laryngeal cancer compared to those with the normal wild-type genotype. There may be no association between the susceptibility to laryngeal carcinoma and the GSTM1-null polymorphism in a population of Romanian descent. Although smoking has been deemed a major carcinogenic risk factor in laryngeal cancer development, no such relationship could be confirmed within this ethnic group for the studied polymorphisms.


We conducted a systematic review and a meta-analysis to assess the risk of serious adverse events in the elderly after yellow fever vaccination compared to the non-elderly population. We searched multiple databases and grey literature, and we selected research without language and publication date restrictions. Studies were analyzed in a descriptive way and meta-analyzed and expressed in terms of prevalence ratio and risk ratio with a 95% confidence interval, depending on the degree of heterogeneity found. A total of 18 studies were included and 11 were meta-analyzed. The results obtained through the meta-analysis showed a risk of serious adverse events after yellow fever vaccination three times higher for the elderly when compared to the non-elderly population and five times higher for persons > 70 years. In relation to adverse event types, viscerotropic disease associated with the yellow fever vaccine had a risk that was six times higher when compared to the population < 60 years. The evidence found supports that the vaccine indication in individuals > 60 years of age should be based on a careful analysis of individual benefit-risk assessments. The results found suggest a higher risk of events for individuals > 70 years, especially for viscerotropic and neurotropic disease associated with YFV contraindicating the use of the YFV in this age group. The yellow fever vaccine (YFV) is the most important means of preventing the disease. It is an immunobiological product considered safe and effective, requiring only one dose to ensure protection against the disease . HoweverIn the last decades, emergency/re-emergence processes of the yellow fever virus have had an important impact on human and animal populations, represented by extensive outbreaks in humans and epizootics in non-human primates, such as that which affected Angola and the Democratic Republic of Congo in 2015 and 2016 ,8,9,10. According to the World Health Organization, by 2020 the number of people over 60 will surpass that of children under 5 years for the first time in history . With thYFV is recommended for people over 60 years of age through a risk-benefit assessment. Due to the phenomenon of immunosenescence, the risk of adverse events occurring in the elderly is greater than that of young adults ,15. The There are systematic reviews that address the analysis of YFV safety in the elderly, however, these included only in articles published until 2012, before the latest major outbreaks of the disease in the world and the increase in vaccine recommendation areas ,17,18,19This systematic review was reported according to the harms checklist from the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA harms) guidelines  Table S. The revClinicalTrials.gov, accessed on 7 November 2021). Those databases were chosen based on the repositories consulted in the previous systematic reviews on the topic and if they were considered reference repositories related to adverse events publication. We used a combination of search terms related to the population, YFV, and outcomes—considering each database’s specific descriptors and free text to increase the search strategy sensitivity as provided in We searched PubMed/Medline, Embase, Web of Science, Scopus, Lilacs, the Database of Abstracts of Review of Effects, and Toxiline as well as sources of grey literature . Study designs considered for inclusion were clinical trials, cohort, case-control, and case series. All different lineages of YFV were considered for inclusion. Studies were excluded if they referred to multiple publications. In that case, only the publication with the most current data was considered.Two independent reviewers (A.A. and L.W.) critically appraised each paper and discussed discrepancies in consensus. We screened titles and abstracts according to specified inclusion and exclusion criteria and then selected the full texts.Data was extracted independently by two reviewers using an electronic form created in Epidata version 4.6. A pilot was conducted with 10% of the studies included in the review.To deal with potential data gaps, authors from the included studies were contacted in up to three attempts by email to obtain unreported or unclear data. If no reply was obtained the study was excluded from the review.The quality of included studies was assessed in accordance with the methodology recommended in the Cochrane Handbook for Systematic Reviews of Interventions using both the Revised Cochrane risk-of-bias tool for randomized trials (RoB 2.0) and the Risk of Bias in Non-randomized Studies-of Interventions (ROBINS-I) tools ,22. Two The GRADE methodology was used to assess the quality of the evidence of the review , using t2 test. The level of heterogeneity was defined as low (I2 = 0% to 25%), moderate (I2 ≥ 25% to 75%), and high (I2 ≥ 75%) [p = 0.05, was also performed. If heterogeneity were above moderate, a narrative analysis summarizing the findings of the studies was conducted using a random effects model [Data extracted from the studies were characterized depending on the homogeneity of the study planning methods and outcomes. An evaluation of heterogeneity was carried out using the I2 ≥ 75%) . The chits model ,25.A meta-analysis was conducted to assess the relative frequencies of SAE in persons > 60 years when compared to those aged under 60 years, using the prevalence rate and 95% confidence interval. Additionally, a meta-analysis of risk factors was conducted for the included studies that presented measures of association, the groups with SAE outcomes were compared with the control group through the risk ratio and the 95% confidence interval.Sensitivity analyses were performed to examine potential explanations of the heterogeneity found between studies by factors related to clinical and methodological characteristics of the studies, such as analysis after removal of outliers, subgroup analysis, and meta-regression. To examine publication bias, the Egger test and the Begg test were performed, as well as an analysis of the funnel graph. The program R version 4.0.2 was used for statistical analysis of the data.We identified 1418 published studies from database searches. No articles were found in the DARE, Open Gray, Gray literature report, Australian Adverse Drug Reactions Bulletin, and FAERS databases.After removing duplicated publications, the first step of selecting articles by reading the titles and abstracts identified 47 papers for full text reading. In this first stage, 14 publications per language were excluded, 6 in Russian, 3 in Arabic and 5 in Chinese.Eighteen studies were identified for analysis ,39,40,41The studies included were published between 2001 and 2020, most with cross sectional design (61.1%) and with a period of data collection greater than 3 years (50.0%). Only three studies were published after the latest yellow fever outbreaks since 2015. Only three studies (16.7%) presented information for a specific age group of the elderly population and 9 (55.6%) did not present information for comorbidities .Regarding the country of conduct of the selected studies, four were multi-country studies ,29,30,32In total, 322,759,087 doses of YFV were administered in the studies evaluated. Of the studies included in the review, eight did not have information to enable the combined analysis of the data and, therefore, were excluded from the meta-analysis ,38,39,412 = 98% and the chi-square (p < 0.01) as well as the visual highlight presented in the forest plot of one study as an outlier [2 = 96% and p < 0.01), there was an increase in the value found for prevalence, with a value of 14 cases for every 1,000,000 vaccinated (CI 95% 0.003–0.068).The samples of the studies included in this review presented data by the number of participants or the distributed/administered doses. Thus, the prevalence was adopted as a frequency measure for the meta-analysis of the 10 studies ,36,37,402 = 60%, p = 0.02) indicates that there are important differences between the results of the studies for this outcome. They identified that the risk was 2.51 times higher for the age group of people ≥ 60 years old (95% CI 1.71–3.69) when compared to the population < 60 years old.Of the ten studies evaluated, seven ,34,35,36To explore possible clinical and methodological factors that could explain the degree of heterogeneity presented in the previous meta-analysis, a subgroup analysis was performed, considering factors related to population, intervention, and study methodology.2 = 30%, p = 0.20) and ≥70 years . Overall, the analysis reveals that the risk of SAE increases with age. The risk of SAE found was 2.32 times higher for <70 years old (95% CI 1.44–3.72) and 4.84 times higher for the age group ≥ 70 years old (95% CI 2.82–8.31) when compared to the population < 60 years old.A subgroup analysis was conducted considering the different age groups of the elderly population to explore the high heterogeneity found in the previous analysis and a potential “dose-response” relationship between the age groups A. The an2 = 44%; p = 0.17) for studies that reported SAE related to viscerotropic disease associated with YFV. For this group, the risk was 6.59 times higher (95% CI 2.14–20.28) for elderlies when compared to the group aged <60 years. For neurological manifestations associated with YFV, although heterogeneity remained high , the risk was 4.45 times higher (95% CI 1.14–17.38).The type of SAE reported was also evaluated in a subgroup analysis B. There 2 = 89%; p < 0.01), which can be explained by their distinct study designs . Other p = 0.0250) and the elderly without age specification (p = 0.0089) showed a greater association with the heterogeneity found when compared to the age group above 70 years. No association was found between heterogeneity and the characteristics assessed for study design and type of SAE.A meta-regression was performed to examine the individual contributions of the studies in the heterogeneity found and the potential associations with age, SAE type, and study design . Age wasp = 0.2637) suggest publication bias, indicating a greater uptake of studies that showed a greater effect related to the risk of SAE. For the risk factor meta-analysis, the symmetry found in the funnel graph and the result found by both tests (p = 0.5961) do not suggest a publication bias.As for the publication bias analysis, for the prevalence meta-analysis both the asymmetry shown in the funnel graph and the result of the Begg test , not receiving the assigned intervention (87.5%), and measuring the intervention (81.2%). The domain with the highest number of studies with a critical bias level was selection (56.2%) . Only onIn general, the findings of this literature review showed a higher risk of SAE for the population over 60 years, and the elderly population had a risk three times higher of developing SAE when compared to the non-elderly population.Furthermore, the results obtained through the meta-analysis showed that this risk is almost five times higher for the age group over 70 years old, when compared to those aged 60 to 69 years. These findings are in line with previous research on this topic ,19,36, aAmong the types of SAE analyzed in the review, the greatest risk was for viscerotropic disease associated with YFV up to six times higher when compared to the population under 60 years. Other studies carried out for this target population found a similar tendency of a higher risk for both viscerotropic diseases associated with YFV and neurotropic disease associated with YFV compared to other types of SAE ,18,38,42Concerning the presence of heterogeneity found, subgroup analysis and meta-regression showed that age is associated with the degree of heterogeneity found in the meta-analysis of risk factors. This may be related to some of the characteristics of the immunosenescence process, in addition to other possible individual characteristics of the elderly <70 years old present in the studies when compared to those aged ≥70 years old ,3,4,5.As for the methodological quality of the studies included in the review, this ranged from moderate to critical depending on the assessment tool used. The most frequently included study designs were non-experimental, such as those that were transversal. This characteristic can also be associated with the high degree of heterogeneity found in the meta-analysis, since the highest risks of bias were related to the selection of participants and losses.The same was reflected in the general analysis of the quality of the review′s evidence that suggests low and moderate confidence in the estimated effect from the included studies. Despite this, it was observed that the evidence evaluated in this study has a strong recommendation against the use of YFV in the target population of the review because it presents a higher risk of SAE, especially in the age group above 70 years.Our study has some limitations. The studies included in this review may have different bias affecting the denominators used, among them the differences between the study designs, the selection of the participants, the loss of information, and the participants, which directly influences the confidence of the conclusions generated in the results found. The same also occurred for the numerators used since the classification of some types of SAE varied between studies.Another limitation is related to the difficulty of investigating SAE for this specific population. This is because the events found may be associated with other pre-existing factors for these individuals, such as comorbidities, which can make it challenging to prove causality between the application of the vaccine and the involvement of the adverse events itself ,19,43. IThe yellow fever vaccine research program has some inherent limitations which may be reflected in the findings of this review. Although we developed a sensitive search strategy and included several database sources, few publications were found on this topic for this specific population. Furthermore, only three studies included were published after the latest outbreaks of the disease, since 2015.This fact can be related to two main factors. First, YFV is not routinely recommended for the target population of this review, but only under medical judgment ,2; seconThus, the evidence found in this review supports the indication of YFV in individuals over 60 years of age must be based on a careful analysis of an individual benefit-risk assessment. However, it is necessary to highlight that the results found suggest a higher risk of SAE for individuals over 70 years, especially for viscerotropic and neurotropic disease associated with YFV contraindicating the use of the YFV in this age group.It is noteworthy that the individual benefit-risk analysis must be carried out with a particular perspective in regions where the epidemiological scenario reinforces a high risk of contracting the disease, such as in endemic regions, where the population is naturally exposed for a longer period. This question should be taken as a basis for users, health professionals, and other decision makers on the use of this technology. Advances in the development of new yellow fever vaccines could be a strategy to avoid serious adverse events. One approach could be inactivated vaccines currently under early-phase clinical studies .Yellow fever is a serious public health concern and without the YFV the risk of an increase in outbreaks and large-scale epidemics would be high, mostly related to the extensive spread of disease vectors and due to an ever-increasing movement of people around the globe. Even though YFV adverse events need to be considered, the vaccine is the best way to avoid the disease, and it should be used in endemic areas and in populations that could potentially be exposed, considering a careful risk-benefit analysis. More research to increase knowledge about the potential risk of SAE in specific contexts for the population studied and to generate new vaccine technologies for the disease, such as inactivated vaccines against YF already underway, is needed to guide the best recommendation for this population.


The association between grip strength (GS) and non-alcoholic fatty liver disease (NAFLD) has been reported by recent epidemiological studies, however, the results of these studies are inconsistent. This meta-analysis was conducted to collect all available data and estimate the risk of NAFLD among people with low GS, as well as the risk of low GS among patients with NAFLD.I2 test, and potential for publication bias was assessed qualitatively by a visual estimate of a funnel plot and quantitatively by calculation of the Begg’s test and the Egger’s test.We systematically searched several literature databases including PubMed, Web of Science, Cochrane Library, and Embase from inception to March 2022. These observational studies reported the risk of NAFLD among people with low GS and/or the risk of low GS among patients with NAFLD. Qualitative and quantitative information was extracted, statistical heterogeneity was assessed using the Of the citations, 10 eligible studies involving 76,676 participants met inclusion criteria. The meta-analysis of seven cross-section studies  showed that people with low GS had increased risk of NAFLD than those with normal GS . In addition, the meta-analysis of four studies  reported that the risk of low GS patients with NAFLD was higher than those in normal people .In this meta-analysis, we demonstrated a strong relationship between low GS and NAFLD. We found an increased risk of NAFLD among people with low GS, and an increased risk of lower GS among NAFLD patients.www.crd.york.ac.uk/prospero], identifier [CRD42022334687].[ Currently, non-alcoholic fatty liver disease (NAFLD) has become one of the most common causes of chronic liver disease, it is defined by the presence of steatosis in more than 5% of hepatocytes with little or no alcohol consumption . NAFLD iIn the United States, approximately 30% of individuals are diagnosed with NAFLD . In ChinGrip strength (GS) is a measure of the maximum static force that a hand can apply around a dynamometer. GS is often considered an indicator of muscle mass and muscle strength . ResearcNon-alcoholic fatty liver disease is a systemic condition that has a bi-directional relationship with the components of metabolic syndrome . AccordiThis meta-analysis followed the Preferred Reports Items for Systematic Reviews and Meta-analyses (PRISMA) reporting guideline . The proTwo investigators (LH and SF) independently conducted an electronic literature search using PubMed, Web of Science, Cochrane Library, and Embase, language was restricted to English, from database inception to March 2022. In PubMed, controlled vocabulary terms and the following keywords were used:  OR  OR  OR  OR  OR  OR  OR  OR  OR  OR  OR  OR  AND (“Hand Strength” [Mesh]) OR  OR (Grip Strength) OR  OR (Hand Grip Strength) OR  OR  OR (Grip) OR (Grips) OR (Grasp) OR (Grasps). A similar search strategy was run in other databases. The database search revealed 224 articles that could have been included in our meta-analysis, and 43 articles were excluded because they were duplicated. After removing duplicates, all titles and abstracts for potential inclusion were screened by two independent researchers (LH and SF). Based on the inclusion and exclusion criteria, 164 articles were excluded after reading the titles and abstracts. Finally, 17 full texts of these records were selected for detailed assessment. The two researchers extracted the related data according to the inclusion criteria. If the studies were potentially eligible for inclusion, the full text was examined. The two reviewers would discuss with each other any disagreements that may have occurred.All studies were assessed for selection and measurement biases according to the Newcastle-Ottawa Scale (NOS) . The NOSThe same two authors evaluated the titles and abstracts of eligible studies and any disagreements were resolved by consensus. The inclusion criteria are as follows: (1) studies on the association between NAFLD and GS; (2) used a standardized index to diagnose and assess NAFLD and GS; (3) reporting odds ratio (OR) and 95% confidence intervals (95% CI) for GS and NAFLD; (4) the full text of the study could be assessed; and (5) the full text of the study could be assessed.in vitro parameters or used animal models; and (6) based on the NOS scores, the low-quality studies were excluded.The same two authors evaluated the titles and abstracts of eligible studies and any disagreements were resolved by consensus. The exclusion criteria are as follows: (1) did not use clear diagnostic criteria for NAFLD; (2) the measurement of GS is not accurate; (3) the study did not provide the OR of NAFLD and GS; (4) case reports, case series, reviews, posters, and abstracts were excluded; (5) measured only Data collection process two independent researchers (LH and SF) assessed the full texts of included studies and used a standard data extraction form when extracting data. Any disagreements were resolved by discussion until consensus was reached. The data extracted for the analysis involved: (1) the first author’s name and publication year; (2) the sample size and number of cases; (3) the mean age and sources of participants; (4) the OR with the corresponding 95% CI; and (5) the scores of NOS in the studies.P-value less than 0.05 was considered to be statistically significant. We collected the summary OR of NAFLD and low GS. The heterogeneity of results between studies was determined by the I2 test , a fixed-effect model was used to pool outcomes, otherwise a random-effect model was applied for the meta-analysis (I2 > 50%). The publication bias was assessed qualitatively by a visual estimate of the funnel plot and quantitatively by calculation of Begg’s test and Egger’s test ], diagnosis of NAFLD [ultrasonography and hepatic steatosis index (HSI)], region (China and Korea), mean age (<60 and >60 years old), several participants . One-study-removed sensitivity analyses were performed to determine the relative impact of each study on the overall risk estimate.Ten articles were included in this analysis  9, 24–3–324–32. The quality level of each study ranged from 5 to 8 stars . The funI2=98%), so we chose the random-effects model. Meta-analysis of these studies showed that low GS patients had odds of NAFLD that were 3.32 times as high as normal GS  and Egger’s test (P = 0.785) suggest that there is no significant publication bias.In the seven studies of the OR of NAFLD included in this meta-analysis , the samBecause of the high heterogeneity, we conducted a series of subgroup analyses to identity the heterogeneity source. The subgroup analysis by several participants revealed no significant difference between numbers , the OR I2 = 65%). Meta-analysis of these studies showed that patients with NAFLD had odds of low GS that were 3.31 times as high as normal group  and Egger’s test (P = 0.630) suggest that there is no significant publication bias.In the four studies of the OR of low GS included in our meta-analysis , the samP > 0.05) showed that age may not be the cause of high heterogeneity. Additionally, we conducted a subgroup analysis according to the method of GS ascertainment (GS and RGS), there were significant association between NAFLD and low GS was detected in the studies using RGS (Because of the moderate heterogeneity, we conduct a subgroup and meta-regression analyses to identify the heterogeneous source. These four studies, all the shown that NAFLD patients have markedly low GS than the non-NAFLD groups. Because meta-regression was performed to examine possible heterogeneous factors for quantitative variables, we used age as a covariate for meta-regression, but the result (In the two analyses, in one-study-removed sensitivity analyses, we excluded each study and results did not change , 13.To the best of our knowledge, this is the first systematic review and meta-analysis that summarized available studies regarding the association between GS and NAFLD. In this meta-analysis which included 10 studies with a total of 76,676 participants, we performed two types of meta-analysis, and explored the OR of NAFLD in patients with low GS and the OR of low GS among patients with NAFLD, both results suggest a significant association between the NAFLD and GS: the Non-alcoholic fatty liver disease encompasses a wide range of diseases from simple steatosis, non-alcoholic steatohepatitis, fibrosis, and even cirrhosis . SkeletaOn the other hand, chronic inflammation is the other hypothesis most often cited. There are several studies focus on the mediators that link the muscle-liver-adipose tissue axis . For exaRecently, low vitamin D levels have been suggested to be associated with NAFLD and muscle strength. Vitamin D plays an important role in muscle mass and muscle strength. A systematic review revealed that vitamin D supplementation significantly increased muscle strength . A separIn addition, gut microbiota is also an important component in the pathological mechanism. The gut microbiota composition is generally shaped in early childhood , and by This study also has several limitations. Firstly, one analysis has moderate heterogeneity, and another analysis has high heterogeneity, which maybe because of the small number of included studies, the restricted regional distribution of the studies, and age differences in each study. Secondly, in the subgroup analyses, some subgroups only have 2 or 3 studies, which may affect the results. Besides, subgroup analyses are observational by nature and may be subject to confounding by study-level characteristics. Finally, the definition of NAFLD is different in included studies, some studies use ultrasound to examine the NAFLD, and some studies use HSI to diagnose NAFLD, which may affect the study findings.Therefore, GS, as an important parameter of sarcopenia and muscle strength, is associated with NAFLD, not only in pathological mechanisms, such as insulin resistance, chronic inflammation, gut microbiota, and regulation of vitamin D, but also in terms of clinical data that people with low GS have a higher risk of NAFLD, and patient with NAFLD have lower GS than normal people.In conclusion, there is an association between NAFLD and GS. Compare with the normal group, people with low GS are more likely to develop NAFLD, in addition, GS levels in NAFLD patients are also generally lower than the normal population.The original contributions presented in this study are included in the article/LH: study idea, concept and design, data extraction and interpretation of data, drafting of the manuscript, and review of the final manuscript. SF: data extraction and analysis of data, drafting of the manuscript, and review of the final manuscript. JL: drafting of the manuscript, data analysis, and review of the final manuscript. DL: drafting of the manuscript and review of the final manuscript. YT: study idea, concept and design, drafting of the manuscript, and review of the final manuscript. All authors contributed to the article and approved the submitted version.


Previous studies showed conflicting results regarding peripheral vitamin D levels in ankylosing spondylitis (AS). We performed this systemic review and meta-analysis to explore whether vitamin D may influence AS process.Articles published until March 2022 were searched in databases as follows: PubMed, Web of Science, and Google Scholar. The present study included cross-sectional and case-control studies regarding vitamin D levels in patients with AS. Studies were excluded according to the following exclusion criteria: (1) we excluded studies which did not provide sufficient information regarding the comparison of vitamin D levels in AS patients and healthy controls (HC). Vitamin D levels in the two group studies should be reported or could be calculated in included studies; (2) meta-analysis, reviews and case reports. STATA 12.0 software was used to make a meta-analysis. Standard mean differences (SMDs) and 95% confidence intervals (CIs) were computed as effect size.The present meta-analysis showed no significant difference in peripheral 1,25-dihydroxyvitamin D3  levels between AS and healthy controls (HCs) in Caucasians with a random effects model . Patients with AS had lower peripheral 25-hydroxyvitamin D (25OHD) levels compared with HC with a random effects model . Patients with AS had higher peripheral C-reactive protein (CRP) and erythrocyte sedimentation rate (ESR) levels compared with HC in Caucasian population with random effects models . However, no significant difference in alkaline phosphatase (ALP), parathyroid hormone (PTH) or calcium levels were indicated between AS and HC in Caucasian with random effects models .In conclusion, the study showed an inverse association between 25OHD and AS, which suggests that vitamin D may have a protective effect on AS. ESR and C-reactive protein (CRP) are important biomarkers for AS. Ankylosing spondylitis (AS) is a chronic immune-mediated progressive rheumatic disease that mainly affects the axial spine, especially, the axial spine and sacroiliac joints . AS is cVitamin D is a fat-soluble that is important in mineral and bone metabolism. The natural way to obtain vitamin D is through safe sunlight exposure and daily diet supplement . VitaminThere are increasing data linking vitamin D and the immune system. Vitamin D deficiency has been described as a risk factor in the progression of several autoimmune diseases such as multiple sclerosis and Crohn's disease, generally attributed to the potent immunomodulatory effects of 1,25-dihydroxyvitamin D3  , 12. KliThe present study was conducted drawing on the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement . The preArticles published until March 2022 were searched in databases as follows: PubMed, Web of Science, and Google Scholar. Search terms (“ankylosing spondylitis” OR “AS”) AND  were used.The present study included cross-sectional and case-control studies regarding vitamin D levels in patients with AS. Studies were excluded according to the following exclusion criteria: (1) we excluded studies which did not provide sufficient information regarding the comparison of vitamin D levels in patients with AS and HC. Vitamin D levels in the two group studies should be reported or could be calculated in included studies; (2) meta-analysis, reviews, and case reports.Q-test and I2 were used to assess heterogeneity among studies. In addition, meta-regression analysis was conducted to detect source of the heterogeneity. With high heterogeneity , random effects models were used to compute results; with low heterogeneity , fixed effects models were used. Subgroup analyses (for different ethnicities) were used to investigate the effect of ethnicities on results. In addition, the stability of the meta-analysis was evaluated with sensitivity analysis. Moreover, the Begg's test, Egger's test, and funnel plots were made to evaluate publication bias.In total, two investigators extracted data from finally included studies independently. Extracted data showed as follows: author, publication year, study location, sample size, gender, mean age, and results. STATA 12.0 software was used to make meta-analysis. Standard mean differences (SMDs) and 95% confidence intervals (CIs) were computed as effect size. Selection procedures were illustrated in p-value of Q test <0.001, I2 = 96.2%; p-value of Q test <0.001, I2 = 94.3%; p-value of Q test <0.001, I2 = 85.0%; p-value of Q test <0.001, I2 = 85.3%; p-value of Q test <0.001, I2 = 95.1%; p-value of Q test <0.001, I2 = 91.7%; p-value of Q test = 0.002, I2 = 76.8%; The present meta-analysis showed no significant difference in peripheral 1,25OHD levels between AS and HC in Caucasian with a random effects model [SMD: −0.68, 95% CI ; Meta-regression analysis showed that ages and gender were not responsible for heterogeneity across studies regarding comparison of peripheral 25OHD, ALP, CRP, ESR, PTH, and calcium levels between patients with AS and HC .Sensitivity analyses showed no changes in the direction of effect when any one study was excluded for all meta-analyses . SupplemIn this systemic review and meta-analysis, we found that there was a significant difference between HC and patients with AS in the level of serum 25OHD, whereas no difference in 1,25OHD. This result was consistent with the article published by Cai et al. in 2015 which reported that the SMD of 25OHD between patients with AS and controls was −0.66 , and 1,25OHD was −0.72  . The conIn total, 1,25OHD and vitamin D are recognized as important immune system regulators that is widely accepted and the vitamin D receptor (VDR) has been found in the immune system . MurdacaPrevious studies have concluded that there is a tightly regulated feedback cycle between vitamin D and PTH secretion . PTH stiWe discussed the associations between activity of AS and several critical factors and found there were significant differences in 25OHD, ESR, and CRP between patients with AS and HC. However, there were some limitations in this meta-analysis. At first, most of included studies were based on Caucasians, and only one on yellow race. Second, because of the limitation of data resources, we could not evaluate the impact of several factors  on the association between AS and vitamin D.In conclusion, the study showed an inverse association between 25OHD and AS, which suggests that vitamin D may have protective effect on AS. However, whether vitamin D supplement decreases the risk of AS needs further research. More well-designed studies are needed to confirm the relationship between vitamin D and AS.The original contributions presented in the study are included in the article/MD planned the study, made data collection, and manuscript writing. JP made study search, data collection, and data analysis. DW made study search and data analysis. HW submitted the study. All authors contributed to the article and approved the submitted version.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.