PSD patients were older  compared with their non-PSD counterparts. Advanced age, atrial fibrillation, prior stroke, coronary artery disease, heart failure, and peripheral vascular disease appeared to be significantly associated with PSD.Vascular risk factors may predispose to post-stroke delirium (PSD). A systematic review and meta-analysis were performed by searching PubMed, Web of Science, and Scopus. The primary outcome was the prevalence of vascular risk factors in PSD vs. non-PSD patients. Odds ratios (ORs) with 95% confidence intervals (CIs) and mean differences (MDs) with 95% CIs were calculated for categorical and continuous variables, respectively. Fixed effects or random effects models were used in case of low- or high-statistical heterogeneity, respectively. We found an increased prevalence of atrial fibrillation (OR = 1.74,  Post-stroke delirium (PSD) comprises a frequent, severe, and yet potentially preventable complication of acute strokes ,2,3. AltRecent meta-analyses have highlighted the burden caused by PSD, and reported that almost 25% of acute stroke patients develop PSD, whereas there are other studies reporting a PSD rate as high as 60% . In termIn clinical practice, currently available PSD treatment options are characterized by limited efficacy ,10. IncrVascular risk factors have been previously linked to PSD ; howeverIn view of this gap, a systematic review of the literature, comprising all of the so-far published literature on vascular risk factors and PSD, was performed. We then aimed to increase the class evidence of the associations between vascular risk factors and PSD through an aggregate data meta-analysis, to assess associations between vascular risk factors and PSD.For the current meta-analysis, we applied the Preferred Reporting Items Systematic Reviews and Meta-Analyses (PRISMA) guidelines (File S1) . This stWe conducted a database search in the PubMed, Web of Science, and Scopus databases for eligible articles presenting data regarding the risk factors for PSD. The last literature search was performed on 28 April 2022. We used the term “stroke” in combination with the term “delirium”. The complete PubMed algorithm of the conducted literature review search for the current meta-analysis was (“stroke”[MeSH Terms] OR “stroke”[All Fields] OR “strokes”[All Fields] OR “stroke’ s”[All Fields]) AND (“delirium”[MeSH Terms] OR “delirium”[All Fields] OR “deliriums”[All Fields]). We identified and removed the duplicates using EndNote citation manager software .The literature was screened for study suitability based on title and abstract. Only subject-relevant studies were therefore considered as eligible. We retrieved the full texts from the articles that passed the initial screening. Furthermore, the reference lists of the included articles (see inclusion criteria below) were screened for further eligible studies that may have been missed from the initial scan.Studies were included in the meta-analysis according to the following criteria:(1) They were published before 28 April 2022; (2) they presented data regarding the risk factors of interest for the two groups (PSD vs. non-PSD) or for the overall sample and at least for one subgroup (PSD or non-PSD); (3) respective data should be presented as absolute numbers or proportions for categorical variables and as means [with the standard deviations (SD)] or median [with the respective range or interquartile range (IQR)] for continuous variables; (4) no limitations were implemented with respect to study design ; (5) delirium detection was either routinely and prospectively performed by attending physicians or via the capitalization of secure, primary hospital records, using standard assessment protocols in both cases; and (6) the studies were written in the English language. In case of multiple articles presenting results from the same sample, data from the article with the largest sample or the one presenting the respective relative data were included. Studies were excluded according to the following criteria:(1) reviews, meta-analyses, and case-reports; (2) studies not involving human subjects; (3) studies with full texts and abstracts not published in English; (4) book chapters, conference abstracts, and posters presentations; (5) studies without (or with infeasible extraction of) relevant data for both groups (PSD vs. non-PSD); (6) studies implementing diagnostic procedures not based on direct (attending physician) or primary medical records assessments -based PSD establishment, self-reported diagnosis, or surrogate-reported diagnosis); and (7) retracted articles.We applied the Newcastle–Ottawa scale (NOS) for case-control and cohort studies and the modified NOS for cross-sectional studies in order to assess the quality of each study, as previously described  24,25].,25.24,25Cochrane Handbook for Systematic Reviews , (6) age, (7) sex distribution, (8) method of PSD assessment, and (9) times of PSD assessments. Moreover, we extracted data regarding histories of hypertension, smoking, alcohol consumption/abuse, previous or current atrial fibrillation (AF), diabetes mellitus (DM), hyperlipidemia, prior stroke , coronary artery disease (CAD) including also myocardial infraction, ischemic heart disease, and atherosclerotic heart disease, heart failure (HF), and peripheral vascular (PVD). The mean value [with the respective SD] or the median value (with either the range or IQR) of the continuous variables (age) and the total number (n) of the categorical variables  were obtained. We applied the method of Hozo et al. (2005), to calculate the mean and SD when the median and range were provided . Moreove Reviews .https://training.cochrane.org/online-learning/core-software/revman/revman-5-download, accessed on 15 December 2021). The odds ratios (ORs) and the respective 95% confidence intervals (CIs) were calculated in order to estimate the effect of the categorical variables  on PSD. We estimated the mean differences (MDs) and their respective 95% CIs for the continuous variable (age). The inverse variance method of individual effects was used. An alpha error of 5% (two-tailed p < 0.05) was established as the statistical significance threshold. In case of an association between a risk factor and an outcome, we performed further subgroup analyses, based on the type of stroke. Two groups were created: 1. Ischemic stroke/TIA and 2. ICH/SAH, whenever possible.Statistical analyses were performed using the Review Manager (RevMan) Version 5.4.1 . After the removal of duplicates, 2137 entries were scanned. Finally, 116 full texts were assessed for eligibility, with 42 articles ,69,70,71Q < 0.00001, I2 = 80%) was revealed. Statistically significant higher age was observed in patients with PSD compared with those without PSD , p < 0.00001) , p < 0.00001] or mixed  type of delirium of the study of Fialho Silva et al. (2021) did not change the results.Thirty studies (8875 subjects) were included in this analysis. For the main analysis, we used the hypoactive (largest sample) type of delirium of the study of Fialho Silva et al. (2021). High heterogeneity (P0.00001) . Sensiti0.00001) . The valQ = 0.07, I2 = 28%). No statistically significant difference for the effect of male sex was observed between patients with PSD and those without PSD , p = 0.79)  were included in this analysis. The analysis revealed high heterogeneity (P = 0.79) . Sensiti = 0.79) .Q = 0.45, I2 = 0%) was observed. There was no statistically significant difference regarding hypertension between patients with PSD and those without PSD   , Alvarez-Perez et al. (2017) , and Hoyosa et al. (2018) , where statistically significant differences were revealed. This can be explained by the fact that these studies reported a lower prevalence of hypertension in the PSD group compared with the non-PSD. The analysis for publication bias revealed no significant asymmetry (Twenty-three studies (7293 subjects) were included in this analysis, while no significant heterogeneity  was observed. There was no statistically significant difference regarding alcohol consumption/abuse between patients with PSD and those without PSD   , where statistically significant differences were revealed. These studies reported a lower prevalence of alcohol consumption at the PSD group compared with the non-PSD. The analysis for publication bias revealed no significant asymmetry, with the study of Ng. et al. (2019) showing a large SE (Fourteen studies (4402 subjects) were included in this analysis, while no significant heterogeneity  was observed. There was no statistically significant difference regarding smoking between patients with PSD and those without PSD  (Fourteen studies (4566 subjects) were included in this analysis, while significant heterogeneity  was observed. There was no statistically significant difference regarding smoking between patients with PSD and those without PSD  (Twenty-six studies (7754 subjects) were included in this analysis. No significant heterogeneity  was observed. There was a statistically significant difference regarding AF between patients with PSD and those without PSD  (Thirteen studies (5016 subjects) were included in this analysis. Significant heterogeneity  was observed. There was no statistically significant difference regarding hyperlipidemia between patients with PSD and those without PSD  (Eight studies (1375 subjects) were included in this analysis. Regarding the study of Shish et al. (2007), we included in the main analysis the lowest estimation sample (meaning the lower number of patients with hypercholesterolemia and/or hypertriglyceridemia) in order to avoid any overlap. Significant heterogeneity  was observed. There was a statistically significant difference regarding the previous cerebral adverse event between patients with PSD and those without PSD  (Twenty-one studies (6670 subjects) were included in this analysis. For the studies of Gustafson et al. (1991), Sheng et al. (2006), Kara et al. (2013), Kotfis et al. (2018), Zaitun et al. (2019), and Pendlebury et al. (2022), we included in the main analysis the lower estimation sample (meaning the lower number of patients with stroke and/or TIA) in order to avoid any overlap. No significant heterogeneity  was observed. There was a statistically significant difference regarding CAD between patients with PSD and those without PSD  (Twelve studies (4966 subjects) were included in this analysis. Regarding the study of Kotfis et al. (2018), we included in the main analysis the lowest estimation sample  in order to avoid any sample overlap. No significant heterogeneity  was observed. There was a statistically significant difference regarding HF between patients with PSD and those without PSD  (p = 0.33] (data not shown). The analysis for publication bias revealed no significant asymmetry (Five studies (1299 subjects) were included in this analysis. No significant heterogeneity  was observed. There was a statistically significant difference regarding PVD between patients with PSD and those without PSD   were included in this analysis. No significant heterogeneity , p < 0.00001], age , HF , and the PVD . The statistical significance was lost for the CAD   were possible for the significantly associated risk factors  for both of the subgroups, with exception of the HF and the PDV for the ICH/SAH analysis. Statistical significance was maintained in the ischemic stroke/TIA group for the six out of seven examined risk factors, namely AF  . Respectp = 0.05], prior stroke , age , and CAD  , = 0.002] . In factIn the present systematic review and meta-analysis, we examined whether vascular risk factors, including age, male sex, smoking, alcohol, hypertension, DM, AF, hyperlipidemia, prior stroke, CAD, HF, and PVD are associated with PSD. Our analyses reveal an increased prevalence of AF, prior stroke, CAD, HF, and PVD in patients with PSD compared with their non-PSD counterparts. These associations retained statistical significance in the subgroup analyses based on stroke subtype (ischemic stroke/TIA and ICH/SAH), with the exception of CAD in the ischemic stroke group and AF in the ICH group. Conversely, hypertension and alcohol consumption did not appear to be associated with PSD, because associations were only evident in a subset of the sensitivity analyses and not in the main analyses. These findings highlight that AF, prior stroke, CAD, HF, and PVD may predispose to PSD, and provide a rationale for the selection of predictors that need to be considered for prospective studies validating a multivariable predictive model for PSD. However, some further considerations should be taken into account for an accurate interpretation of the present findings.In our analysis, AF was almost two-fold more prevalent in patients with PSD. This is of particular interest, considering that AF is the most common cardiac arrhythmia  and thatBased on our meta-analyses, the prevalence of CAD was almost 1.5-fold higher in PSD patients. The frequency of delirium among patients with CAD appears to be lower compared with those with stroke ,79,80. COur analyses revealed that HF and PVD were two-times more frequent in PSD patients. However, these results derived from a relatively small number of studies . Moreover, the ORs follow that of the most-weighted study , the exclusion of which in the sensitivity analysis led to no-longer statistically significant results. Finally, as HF and PVD have also been reported as risk factors for delirium in non-stroke patients ,84, furtWe also found that patients with PSD were 5.27 years older compared with non-PSD patients. Age is a non-modifiable risk factors, and there is a great amount of evidence of its contribution to delirium . MoreovePrior stroke was associated with the risk of developing PSD. This may be explained by the fact that prior stroke may lead to greater clinical impairment and thus to increased complications (such as aspiration pneumonia or immobilization), which could also lead to delirium ,50,87. There are additional factors, apart from those examined in our meta-analysis, that possibly contribute to the development of PSD. Pre-existing cognitive decline/dementia have been reported to attribute an almost four-fold higher risk for PSD . MoreoveA better understanding of the predisposing factors leading to PSD could be beneficial for the stroke’s overall cost-of-illness. Indeed, PSD is related to a greater overall cost of hospitalization, though without being attributed to a single factor . MoreoveAmong the strengths of our study is the extended literature search, obtaining a comprehensive summary of the published literature evaluating PSD and risk factors. Moreover, we performed sensitivity analyses, providing additional robustness to our conclusions. At this point, the limitations of this meta-analysis must also be acknowledged. Firstly, in an attempt to maximize the study population, we included studies irrespective of their design, the time of delirium assessment, and the delirium type. Moreover, there was heterogeneity in the included examined phenotypes, as few studies presented merged data by combining different stroke sub-phenotypes. Thus, in the main analysis we present the effects of the risk factors in a combined phenotype consisting of patients with ischemic stroke, ICH, TIA and SAH, with additional subgroup analyses based on the type of stroke in case of a significant association of the examined risk factor with PSD. Moving on, we focused only on predisposing vascular risk factors and stroke severity, as we did not meta-analyze other potential co-factors . Furthermore, we meta-analyzed aggregate data and, thus, the reported associations cannot be adjusted for subjects’ baseline characteristics and potential PSD confounders. Finally, though unlikely, the possibility that some eligible studies failed to be identified cannot be completely ruled out. In conclusion, with the current meta-analysis, we provide quantitative data that suggest the increased prevalence and possible causal-contributions of age, AF, prior stroke, CAD, HF, and PVD to PSD development. Considering the as-of-yet limited and often ineffective therapeutic approaches for PSD, it will be of interest to observe to what extent a personalized prevention and treatment of the modifiable PSD vascular risk factors could be effectively applied to lower the occurrence of PSD. Towards this direction, further collaborative and multiethnic studies are needed in order for PSD to be more effectively prevented and managed.


Localized Gleason Grade Group 5 (GG5) prostate cancer has a poor prognosis and is associated with a higher risk of treatment failure, metastases, and death. Treatment intensification with the addition of a brachytherapy (BT) boost to external beam radiation (EBRT) maximizes local control, which may translate into improved survival outcomes.A systematic review and meta-analysis was performed to compare survival outcomes for Gleason GG5 patients treated with androgen deprivation therapy (ADT) and either EBRT or EBRT + BT. The MEDLINE (PubMed), EMBASE and Cochrane databases were searched to identify relevant studies. Survival probabilities for distant metastasis-free survival (DMFS), prostate cancer-specific survival (PCSS), and overall survival (OS) were extracted and pooled to create a summary survival curve for each treatment modality, which were then compared at fixed points in time. An additional analysis was performed among studies directly comparing EBRT and EBRT + BT using a random-effects model.Eight retrospective studies were selected for inclusion, representing a total of 1393 EBRT patients and 877 EBRT + BT patients. EBRT + BT was associated with higher DMFS starting at 6 years  and extending out to 10 years , with an overall hazard ratio of 0.53 (p = 0.02). There was no difference in PCSS or OS between treatment modalities. Differences in toxicity were not assessed. There was a wide range of heterogeneity between studies.The addition of BT boost is associated with improved long-term DMFS in Gleason GG5 prostate cancer, but its impact on PCSS and OS remains unclear. These results may be confounded by the heterogeneity across study populations with concern for a risk of bias. Therefore, prospective studies are necessary to further elucidate the survival advantage associated with BT boost, which must ultimately be weighed against the toxicity-related implications of this treatment strategy. However, approximately 15 % of patients with localized prostate cancer have high or very high risk disease, which carries an increased risk of recurrence following definitive treatment and subsequent progression to cancer-related death In particular, Gleason Grade Group 5 (GG5) disease is an independent predictor of a worse prognosis and denotes high or very high risk disease even when other disease characteristics are favorable When treating high risk disease with external beam radiotherapy (EBRT), different methods of treatment intensification are currently available, including the addition of ADT, extending the duration of ADT, and a brachytherapy (BT) boost. BT boost has already been shown to improve biochemical and local disease control in patients with intermediate and high risk disease, though its benefit in the Gleason GG5 subgroup is unclear 22.1The Preferred Reporting Items for Systematic reviews and meta-Analyses (PRISMA) 2020 statement was used as guideline for this review. A search strategy was first developed using the population, intervention, control, outcome, study design (PICOS) framework . MEDLINEAfter the exclusion of duplicate records, a total of 982 unique results were obtained from MEDLINE (PubMed) and EMBASE and Cochrane. The initial screening of manuscript abstracts, with a brief assessment of full text if necessary, was completed by two authors (SG and TT). Any disagreements were resolved by a third author (LM). Articles were excluded if data for Gleason GG5 disease was not reported; radiotherapy was not the primary treatment modality; patients had recurrent or metastatic disease; or primary data was not reported, as in population database studies. Case reports, case series and articles not in English were also excluded. An in-depth full-text review of 46 manuscripts was then conducted, with additional exclusions made on the basis of: lack of separately reported outcomes for Gleason GG5 disease; overlapping patient cohorts, in which case the broader publication was included; lack of survival data; and total number of Gleason GG5 patients <10. Again, this was initially performed by two authors (SG and TT), with a third author (LM) resolving any discrepancies. The PRISMA flow diagram in 2.2T-stage at diagnosis, use of pre-treatment staging, median ADT duration, treatment volumes, brachytherapy isotopes and technique, median prostate dose and median number of fractions. In studies that did not separately report baseline characteristics for Gleason GG5 patients, these baseline characteristics were assumed to be similar to those of the overall cohort. Assuming an ⍺/β ratio of 1.5, the equivalent dose in 2-Gy fractions (EQD2Gy) was calculated for the EBRT and BT treatments. The total radiation dose to the prostate for patients receiving EBRT + BT was calculated by summing the EQD2Gy values for the individual EBRT and BT treatments.Data extraction was performed by two authors (SG and TT). Study characteristics were identified, including study type, study design, publication date, enrollment period and sample size. Baseline patient characteristics were also extracted, including median prostate specific antigen (PSA) at diagnosis, clinical Survival probabilities were extracted from the Kaplan-Meier curves for distant metastasis-free survival (DMFS), prostate cancer-specific survival (PCSS) and overall survival (OS) in the included studies, where available. Where mortality functions were reported, the complementary survival probabilities were calculated. In studies reporting survival data amongst specific subgroups, the survival probabilities for each subgroup were extracted. Wherever possible, the number of patients at risk at each time interval were extracted directly from the published studies. In cases where numbers at risk were not reported, these values were estimated using an established method that accounts for censoring 2.3The modified Oxford Centre for Evidence-Based Medicine Levels of Evidence criteria were used to assess the quality of each of the included studies 2.4Baseline patient and treatment characteristics were summarized for each treatment group using a weighted average of reported medians based on sample size. Survival probabilities were pooled across studies to produce a summary survival curve using a modification of the random-effects meta-analysis method proposed by DerSimonian and Laird 33.1A total of eight studies were selected for inclusion in the meta-analysis, all of which were retrospective series classified as Level 4 evidence 3.2The baseline patient and treatment characteristics for Gleason GG5 patients from all included studies are summarized in The proportion of patients receiving pelvic nodal irradiation was 47 % and 41 % in the EBRT and EBRT + BT groups, respectively. However, the use of nodal irradiation within the EBRT + BT group may be underestimated, as pelvic lymph node coverage was left to the discretion of the treating physician in Tilki et al. and exact numbers were not reported 3.3Four studies reported DMFS rates after EBRT and five after EBRT + BT. 2 = 90 % . Heterogeneity was high with I2 = 90 % A.Fig. 3F3.4Four studies reported PCSS rates after EBRT and five after EBRT + BT. 2 = 75 % . Heterogeneity was high with I2 = 75 % B.3.5Four studies reported OS rates after EBRT and five after EBRT + BT. Two studies directly compared OS after EBRT and EBRT + BT. There was a trend toward improved OS with BT boost , but this was heavily weighted towards Kishan et al. with only a minor contribution from Yamazaki et al. (not shown) 4This meta-analysis of eight retrospective studies suggested a DMFS benefit with the addition of BT boost in the treatment of Gleason GG5 disease. This was true in both the pooled survival curve and pooled HR analyses. Differences in DMFS were first observed at six years post-treatment and extended out to ten years. This finding is consistent with other studies evaluating the effect of BT boost in Gleason GG5 prostate cancer. In a multi-institutional retrospective series of approximately 1000 patients, Foster et al. identified an increase in freedom from distant metastases with the addition of BT boost, which was most pronounced in the subset of Gleason GG5 patients with additional poor-prognosis risk factors In localized prostate cancer, DMFS has been shown to be a surrogate endpoint for overall survival As far as the rationale behind how BT boost could reduce the rate of metastasis, it has been hypothesized that by minimizing the risk of local failure, BT boost can limit the risk of a subsequent wave of systemic spread T-stages, there is retrospective evidence to suggest that BT boost is still important in T3b-T4 disease The current American Society of Clinical Oncology (ASCO)/Cancer Care Ontario (CCO) guidelines support the use of BT boost in high risk prostate cancer As for the relevance of age, the overall median age of patients in this meta-analysis was approximately 70, suggesting that patients in this age range still stand to benefit from BT boost. A recent retrospective study demonstrated patients above the age of 70 treated with BT boost did not have worse oncologic outcomes compared to their younger counterparts It is also important to contextualize the potential benefits of BT boost with the risk of additional treatment-related toxicity. In the periprocedural period, there is a risk of BT-associated complications like urinary retention requiring prolonged catheterization, which can have a short-term impact on patients’ quality of life. The risk of such complications, particularly urinary retention, is often related to specific baseline patient characteristics As the first meta-analysis of its kind to date, this study provides further support for the use of BT boost in the radiation treatment of Gleason GG5 disease. These findings are applicable to the modern era of radiotherapy as all included studies were published after the year 2015 and utilized long-term ADT. However, this meta-analysis does have several limitations. All included studies were retrospective in nature, as there were no prospective series or randomized trials reporting separate oncological outcomes for Gleason GG5 disease in this context. Retrospective studies have inherent biases, including selection bias, which may confound the observed differences between EBRT + BT and EBRT. Additionally, Kishan et al. alone contributed more than half of the patients in both the EBRT and EBRT + BT groups 5In conclusion, this meta-analysis provides evidence that the addition of BT boost to EBRT and ADT improves DMFS in Gleason GG5 prostate cancer, but its effect on PCSS and OS remains unclear. However, these results may be confounded by the heterogeneous study populations and the overall risk of bias across studies. As this meta-analysis did not compare toxicity outcomes between treatment modalities, it remains unclear if BT boost results in additional toxicity. Ultimately, given the lack of high-level evidence available, prospective trials are needed to further evaluate the survival benefits of this treatment strategy and contextualize them against the toxicity-related implications of BT boost.None.The authors declare that they have no known competing financial interests or personal relationships that could have appeared to influence the work reported in this paper.


The use of hair dye for cosmetic purposes appears to be increasing worldwide. As 50–80% of women use hair dye throughout their lifetimes, the possible association between hair dye use and cancer is a public health concern.This systematic review was performed by retrieving studies from PubMed, Scopus, WOS, and ProQuest databases. The inclusion criteria were case-control studies evaluating the association between hair dye use and cancer in women. Women with cancer who have used any hair dye were the focus of our study.The present study combined 28 studies, to assess the association between hair dye use and cancer. The pooled odds ratio (OR) of hematopoietic system cancers among those who have generally ever used any type of hair dyes was 1.10 (95% CI:1.01–1.20) in 17 studies. In 11 studies investigating hair dye made before and after 1980 as a risk factor for cancer, the pooled OR for cancer was 1.31(95% CI:1.08–1.59). Likewise, in the 13 studies that evaluated the association of light and dark hair dye with cancer, the risk among those using dark hair dye increased by 9%, compared to non-users .The present study suggests that, although the use of hair dye may increase the risk of cancer among users, a more detailed evaluation is required to assess the type of hair dye use in terms of guidelines and metrics. Hair dye is a commercial product that numerous people, particularly women, use to change their appearance, including covering their grey hairRegarding the controversial statements of available related studies and the uncertainty on the risk of cancer from using hair dyes, a systematic review and meta-analysis were conducted to investigate the various reports on the effects of hair dye exposure. Various types of cancer, the effects of the color intensity, and other significant factors were investigated.Protocol. The protocol was submitted to Prospero under the registration code CRD42017074393. This systematic review was performed based on PRISMA guidelinesEligibility criteria. All case-control studies, including population-based and hospital-based case-control studies, were selected. Based on the primary outcome, only studies that evaluated any cancer were included, whereas other outcomes, such as any Medical Syndromes and Cardiovascular diseases, were not considered. Other studies, such as case series and case reports, were excluded. Studies in men were excluded, and when studies included both men and women participants, only data from women were extracted for meta-analysis. Furthermore, one article published in the other language was excluded.Information source. A search was conducted through four databases: PubMed, Scopus, ProQuest, and Web of Science, from January 1990 through 30 July 2017. The search strategy included the keywords ‘cancer’ AND ‘hair dye’. Two study investigators performed the search independently (MA and ME) and were checked by a third investigator (RAN).Search strategy. These search terms were used to perform a final search through the PubMed database as follows:(‘hair dye’[tiab] OR (dyes[tiab] AND hair[tiab]) OR ‘hair colorants’[tiab] OR (colorants[tiab] AND hair[tiab]) OR (‘coloring agents’[tiab] AND hair[tiab]) OR (agents[tiab] AND ‘hair coloring’[tiab]) OR ‘hair coloring agents’[tiab]) AND (cancer[tiab] OR neoplasia[tiab] OR neoplasias[tiab] OR neoplasm[tiab] OR tumor[tiab] OR ‘benign neoplasm’[tiab] OR (neoplasm[tiab] AND benign[tiab]) OR malignancy[tiab] OR cancer[tiab]) AND (1990/01/01:2017/7/31[dp]).This phase was performed in 3 steps. First, duplicate publications were eliminated. Subsequently, each study was screened independently for eligibility criteria (ME and MA); disagreements were resolved by a subject expert (RAN). Finally, the studies screened in the previous steps were selected based on full-text assessment independently by two reviewers (RAN and AHO).To evaluate the methodological quality of the selected studies, the Newcastle-Ottawa Scale (NOS) tool was usedExtraction of data was performed by two reviewers independently (RAN and MA), while >90% of the items were in agreement and other items were resolved by consensus. The extraction data form included general information, quality assessment score, population description, and other related variables. The statistical section of this form categorized OR according to the type of use, type of color, duration of use, and use before/after 1980.Data were analyzed by the STATA 11 software (StataCorp LP). Heterogeneity was assessed by the I2 index and Q test. The random-effects model was used based on the heterogeneity index for pooled estimation, and funnel and forest plots were drawn. A value of 0% indicates no observed heterogeneity, and larger values show increasing heterogeneity.By searching the databases, 1,201 articles were initially obtained; after removing duplicate publications, 428 articles remained.A total of 166 articles were identified after the screening of the title and abstract. Finally, 32 articles were selected following a full-text review. The study selection process is presented as the PRISMA flowchart in With the use of NOS as an assessment tool of case-control study quality, 16/31 studies were of good quality , 40, 41,Analysis of results. The meta-analysis combined 28 studies  to assess the association between hair dye use and cancer. It was demonstrated that the overall cancer risk was increased by 10% in the subjects compared with controls)OR=1.10; 95% CI: 1.01–1.21), and the heterogeneity between studies was moderate (I2=58.2%). There was no evidence of publication bias by funnel plot analysis (data not shown). In addition, subjects who regularly use hair dye had a 10% higher risk of hematopoietic system cancer compared with non-users)OR=1.10, 95% CI: 1.02–1.20) and the incidence of solid cancers was increased by 9% . There wIn addition, of the 28 studies included in the meta-analysis, 11 studies investigated hair dye manufactured before and after 1980 as a risk factor for cancer. In the present study, it was observed using hair dye manufactured before 1980 was associated with increased cancer rates by 31% compared with subjects who did not use hair dye) OR=1.31, 95% CI: 1.08–1.59) , with moAccording to the results of the pooled OR from the 13 studies that evaluated the effect of light and dark hair dye of cancer, the risk on cancer in women using dark hair dye increased by 9% in comparison with women who did not use hair dye  , and theAs indicated by the pooled OR from 18 studies that investigated the association of hair dye stability with cancer, subjects who used permanent hair dye had a 15% higher risk of developing cancer compared with those who did not use hair dye  , and hetBased on the 5 studies that were conducted to investigate the role of duration of hair dye use as a risk factor for cancer, there was no difference in the risk of developing cancer between subjects who used hair dye for >10 years  and thosThe present systematic review and meta-analysis evaluated 28 studies published between January 1990 and July 2017. The analysis revealed a 10% increased risk of cancer by ever use of any type of hair dye. Our results were according to production date (before or after 1980), duration of use, type of dye (permanent or non-permanent), and kind of color (light or dark).Takkouche et al.In two meta-analysis studies conducted by Klesh et al. 2008 While another meta-analysis cancer related to ever use of any type of hair dye, Towel et al.Comparison between light and dark hair dye revealed that the risk of cancer among dark hair dye users was 9% higher compared with that in the control group and that there was a significant association between light hair dye use and cancer. In a meta-analysis of case-control and cohort studies by Klesh et al.These results indicate a significant association between cancer and the duration of hair dye use. The risk of cancer in subjects using hair dye for >10 years was 14% higher than that in the control group.In two meta-analyses based on case-control studies by Klesh et al.In a systematic-review and meta-analysisNitrosamines, produced from secondary amines, are significant components of oxidative hair dyesThe strength of our analysis is the inclusion of all the studies from four databases examine the association between hair dyes and cancer, with a total of >12,313 cases, allowing us to investigate some of the critical variables in the assessment of exposure. However, there were certain limitations to the present study. A methodological limitation was the type of studies included in this evaluation. Selection bias is a particular problem inherent to case-control studies, leading to non-comparability between cases and controls. These studies are also prone to other forms of bias, such as recall and observer bias. If there is dissimilar reporting between cases and controls, the study would be prone to recall bias. If the cases were more likely to report hair dye exposure, the actual effect might be misestimated.The present study aimed to conduct a systematic and meta-analytical assessment of the correlation between hair dye exposure and cancer. The results confirm this association, consistently with those reported by Gera et al.


Flavonoids seem to have hormone-like and anti-hormone properties so that the consumption of flavonoids may have potential effects on hormone-related cancers (HRCs), but the findings have been inconsistent so far. This meta-analysis was aimed to explore the association between flavonoids intake and HRCs risk among observational studies.Qualified articles, published on PubMed, EMBASE, and China National Knowledge Infrastructure (CNKI) from January 1999 to March 2022 and focused on relationships between flavonoids  and HRCs , were retrieved for pooled analysis. Random effects models were performed to calculate the pooled odds ratios (ORs) and corresponding 95% confidence intervals (CIs). Funnel plots and Begg’s/Egger’s test were used to evaluate the publication bias. Subgroup analyses and sensitivity analyses were conducted to explore the origins of heterogeneity.All included studies were rated as medium or high quality. Higher consumption of flavonols , flavones  and isoflavones  was associated with a decreased risk of women-specific cancers , while the higher intake of total flavonoids was linked to a significantly elevated risk of prostate cancer . A little evidence implied that thyroid cancer risk was augmented with the higher intake of flavones  and flavanones .The present study suggests evidence that intake of total flavonoids, flavonols, flavones, flavanones, flavan-3-ols and isoflavones would be associated with a lower or higher risk of HRCs, which perhaps provides guidance for diet guidelines to a certain extent.CRD42020200720.This protocol has been registered on PROSPERO with registration number The online version contains supplementary material available at 10.1186/s12937-022-00778-w. Hormone-related cancers (HRCs) are greatly influenced by hormone levels and generally respond to hormone regulation, which plays an indispensable role in tumor growth. Six cancer types, including breast, ovarian, endometrial, thyroid, prostate and testicular cancer, are usually referred to as HRCs since they share the same carcinogenic mechanism . WorldwiFlavonoids represent one of the largest groups of phytochemicals, as a class of polyphenols with a basic benzo-γ-pyrone structure widely distributed in the plant kingdom, which mainly exist in leaves, flowers, roots, stems, and fruits of plants . The U.SMost of these studies on flavonoids and HRCs have been done in vivo or in vitro, but less in humans. The evidence for the relationship between intake of flavonoids and risk of the aforementioned HRCs (except testicular cancer) among observational studies has been summed up by several existing meta-analyses –24. It iAs of the last summary of evidence, new research has come out, and an outline of the effects of total, subclass of, and individual flavonoids intake on HRCs has so far not been available. Thus, in this study, a meta-analysis was carried out for a comprehensive evaluation of total/subclass of/individual flavonoids to summarize epidemiological findings and to elucidate the association between flavonoids intake and HRCs risk.http://www.ncbi.nlm.nih.gov/pubmed/) and EMBASE (http://www.embase.com/) databases for English studies and on China National Knowledge Infrastructure  for Chinese articles published through January 1999 up to March 2022 was performed. The search terms used for retrieval were as follows: ((“flavonoids”[MeSH Terms] OR “flavonoid$”[All Fields]) OR (“flavanones”[MeSH Terms] OR “flavanone$”[All Fields]) OR  OR (“flavones”[MeSH Terms] OR “flavone$”[All Fields]) OR (“flavan-3-ol$”[All Fields] OR “flavanol$”[All Fields]) OR (“flavonols”[MeSH Terms] OR “flavonol$”[All Fields]) OR “anthocyanidins”[all fields]) AND  OR “endometrial”[all fields] OR “prostate”[all fields] OR  OR “thyroid”[all fields]) AND (“neoplasms”[MeSH Terms] OR “neoplasms”[All Fields] OR “cancer”[All Fields]) AND (“Epidemiologic Studies”[MeSH Terms]) AND (“1999/1/1”[Date—Publication]: “2022/3/31”[Date—Publication]). Simultaneously, we inspected references cited in retrieved articles to recognize extra studies and undertook to get access to unpublished data. The meta-analysis was conducted following the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines. This protocol has been registered on PROSPERO with registration number CRD42020200720 (https://www.crd.york.ac.uk/prospero/).A systematic search on PubMed  an observational study ; (ii) the association between intake of total/subclass of/individual flavonoids and the risk of any HRCs was evaluated; (iii) relative risks (RRs), odds ratios (ORs) or hazard ratios (HRs) with the corresponding 95% confidence intervals (CIs) for the highest category of exposure were reported; (iv) evaluating flavonoids consumption by dietary questionnaires and measurement for serum, plasma and urine. Only the research with the longest follow-up period was taken into consideration when multiple articles involving the same cohort have been produced.The exclusion criteria were as described below: (i) out of adequate statistics; (ii) incomplete evaluation on flavonoids ; (iii) unconfirmed comparison and comparison by per each increase for flavonoids intake; (iv) reported as an abstract, summary, comments, review or editorial; (v) flavonoids intake from sources other than dietary questionnaires, serum, plasma and urine; (vi) amount of soy protein or soy foods or soy supplements consumed as a representative of dietary isoflavone intake. The selection process and the examination for retrieved articles were performed independently by two authors (F. Liu and Y. Peng). Inconsistencies were referred to another author (Yating Qiao) for resolution.Two authors (F. Liu and Y. Peng) reviewed full-text articles separately, and extracted the following characteristics from the original: first author’s name, publication year, study design , study region (country or area), sex and age range, number of cases and controls or total cohort participants with follow-up period, exposure measurement, the most adjusted ORs in case–control studies or HRs/RRs in cohort studies and the corresponding 95% CIs comparing the highest vs. lowest (or reference) level of flavonoids intake, and the covariates in multivariable models. Dissents on data extraction were settled through argumentum.The Newcastle–Ottawa Scale (NOS) quality assessment criteria, consisting of a total of 9 points in three dimensions , was employed to appraise the quality of selected study . StudiesI2 was applied for the evaluation of heterogeneity among studies [p values based on the Q test of heterogeneity were reported (p < 0.05 suggested significant heterogeneity). Publication biases were evaluated by funnel plots, and calculated by Begg’s tests [Statistical analyses were performed by using Stata 15.1 software . ORs with 95% CIs were assessed for ascertaining the association between flavonoids intake and HRCs risk. The merged ORs with 95% CIs calculated by the most adjusted HRs, RRs or ORs with 95% CIs announced in the original were illustrated in forest plots using random effects models depicted by DerSimonian-Laird method . Statist studies  and the ’s tests  and Egge’s tests . BesidesA flowchart of the literature screening process was outlined in Fig. I2 = 76.7%; p < 0.001) and flavones , compared with that in those with lower consumption  –50 were ion Fig. .Fig. 2FoI2 = 0.0%; p = 0.340), and isoflavones  respectively, but not with higher consumption of total flavonoids and other flavonoid subclasses. And, subgroup analysis suggested isoflavones might be more phylactic against ovarian cancer in case–control studies and in Asian populations . Whereas, there was a significant association of isoflavone consumption with reduced risk of endometrial cancer  and flavanones  with no evidence of heterogeneity, respectively  –68 have ely Fig. .I2 = 0%; p = 0.484), which was more pronounced in cohort studies. And, this positive relationship was found only in non-Asia populations , flavones  and isoflavones  was associated with a decreased risk of women-specific cancers, especially among the case–control studies  and 20 case–control studies on breast, endometrial and ovarian cancer were selected for the analysis on the flavonoid intake with the risk of women-specific cancers. No evidence on association between total flavonoids intake and women-specific cancer risk was found in our analysis. For flavonoid subclasses, higher consumption of flavonols  was associated with significant increased risk of men-specific cancers including prostate and testicular cancer, which was restricted to prospective studies  and women-specific cancers .As shown in Fig. Sensitivity analyses for total flavonoids and HRCs were performed Fig. . When anIn the present study, a comprehensive meta-analysis of existing observational studies on the consumption of flavonoids and risk of HRCs were conducted. The current results showed that higher consumption of total flavonoids was only associated with an increased risk of men-specific cancers, mainly prostate cancer. For the subclasses, flavonols, flavones, and isoflavones, as well as the three main individual compounds of isoflavones  may have protective effects on women-specific cancers, whereas potentially dangerous effects of flavones and flavanones have been found in thyroid cancer. There was no evidence in support of any role for anthocyanidins in HRCs.Flavonoids can interfere with the initiation, promotion and development of cancer through directly inhibiting the process of oxidative stress and oxidative damage  and indiConsistent with the findings herein, Chang et al.  implies Menopausal status as an effect modifier may have an influence on the association between flavonoids intake and women-specific cancer risk. Since most studies reported menopausal status limited to breast cancer, the present study conducted subgroup analyses just for breast cancer. A significant association of higher flavones intake in premenopausal women and of higher flavan-3-ols intake in postmenopausal women with reduced risk of breast cancer was found in this study. Isoflavones intake has been reported to be negatively associated with the risk of breast cancer in postmenopausal women , 19, butDietary flavonoids are widely present in plant-based foods. Flavonols, the most abundant flavonoids in foods, are mainly found in onion, red wine, olive oil, berries, and grapefruit. The primary sources of flavones, flavanones, flavan-3-ols and anthocyanidins are vegetables, citrus fruits, green tea and colored berries, respectively. Isoflavones are present in significant quantities in legumes mainly soybeans, green beans, mung beans. The bean of the soy plant has been used in both foods and phytotherapeutic supplements. Evidence from pre-clinical studies supports a role for soy/soy isoflavone in reducing the risk of developing prostate cancer and inhibiting its progression . TherefoDue to the wide varieties of available flavonoids, it is tough to assess the consumption of total flavonoids or each kind of flavonoid. The difficulty of measuring flavonoids intake and the limited epidemiological studies made our research to a certain degree of heterogeneity. Most studies used a food frequency questionnaire (FFQ) or a dietary history questionnaire (DHQ) to evaluate flavonoid consumption, which is likely to lead to measurement bias; while other studies assessed exposure on urine or serum levels, but just reflected the short-term exposure of subjects, in consequence, randomness cannot be ignored. Besides, in addition to studies on morbidity, we also included some cohort studies with follow-up outcomes of recurrence or death to jointly probe the association between flavonoids and HRCs, which is presumably one of the origins of heterogeneity.The research has some advantages that deserve attention. First, this meta-analysis systematically summarized the effective evidence between flavonoids and HRCs. Second, we objectively launched a comprehensive search for studies from 1999 to 2022. Compared with previous meta-analyses, this study included more cohort studies and case–control studies, which undoubtedly provides more tangible and favorable evidence for guiding flavonoids in the prevention of HRCs. Third, most categories of flavonoids were evaluated, including total flavonoids, flavonoid subclasses, and individual compounds in each subclass. Moreover, inclusion and exclusion criteria established were relatively strict. All individual studies included were rated as medium or high quality, which somewhat avoided the insufficiency and unreliability of the results conferred by low-quality studies.On the contrary, there are some limitations and challenges to be settled in the research. As only observational studies were included, evidence power is not very strong yet. The sample size of individual studies is small and there may still be other unknown confounding factors affecting the results. This limits the opportunities for extrapolation. Further intervention studies are needed to confirm our conclusion. The number of eligible studies is still limited, especially for certain cancers, thus flavonoid subclasses cannot be synthetically analyzed. Further, the criteria for the included studies allowed no subgroup analyses based on dietary and other sources (urine/serum/plasma) of flavonoids to be performed. For the sake of avoiding unnecessary heterogeneity, RCTs were not included in our meta-analysis. Limited clinical trials have prevented our results from providing more conclusive evidence.In conclusion, there is a small amount of evidence that total flavonoids, flavonols, flavones, flavanones, flavan-3-ols and isoflavones may be associated with a lower or higher risk of certain HRCs, which maybe provide guidance for diet guidelines to a certain extent in the future. Nonetheless, due to limited data, further prospective studies and a larger number of subjects are warranted to be verified and provide stronger evidence.Additional file1.


Few studies have been conducted to assess the financial balance of the use of genetic signatures in breast cancer, as was published in this journal by Myburgh  with MaBoth studies have limitations such as exchange rate variation, common in developing countries, which alters the costs of chemotherapy and genetic platform; also, because the indirect costs of the treatment were not considered, which would probably further increase the resource savings determined by the application of MammaPrint™. It is worth emphasizing the behavior of MammaPrint™ in different populations, which can determine different outcomes with the application of the test .The analysis of the studies by Myburgh (1) and Mansani (2) show that the use of MammaPrint in the population at high clinical risk, in developing countries, results in significant savings in resources, which are often limited in these regions.I declare that receive funding from Agendia and GenCell Pharma for researching MammaPrint™ in the brazilian population.


Echinococcosis is a severe zoonotic disease that imposes a substantial burden on human life. This meta-analysis aimed to summarize available data on the prevalence of human echinococcosis and identify the key risk factors for echinococcosis in the Chinese general population.2 and Q statistics were calculated to evaluate the heterogeneity, and potential sources of heterogeneity were identified using sensitivity analysis and subgroup analysis. Publication bias was estimated by funnel plots and Egger's test.Relevant studies were comprehensively searched in the PubMed, EMBASE, Web of Science, Cochrane, Chinese National Knowledge Infrastructure (CNKI), Chongqing VIP Information (VIP), Wanfang and SinoMed databases until August 22, 2020. A random-effects model was used to estimate the pooled odds ratio (OR) and 95% confidence interval (95% CI). The I2 = 0), herdsman occupation , not washing hands before meals  and being female  were risk factors for AE. The top five risk factors for CE were ethnicity , nomadism , drinking nonboiled water , feeding viscera to dogs , and herdsman occupation .A total of 1026 studies were identified through the database search, of which 26 were eligible for this meta-analysis. The pooled prevalence of AE and CE were 2.88% and 5.66%, respectively. Ethnicity  in China over the years. The results support that the ethnicity and dog-related factors are major risk factors for both CE and AE. The identification of echinococcosis risk factors may aid researchers and policymakers in improving surveillance and preventive measures aimed at reducing  Echinococcus granulosus and Echinococcus multilocularis, respectively   and alveolar echinococcosis (AE) are the two most common forms of human echinococcosis and are caused by the larval stages of ectively . Dogs aranulosus , whereasocularis . Both ar (DALYs) ; It is aE. granulosus, the potential risk factors for E. multilocularis are more complex because its life cycle involves multiple wild canids as final hosts and a large number of small mammals (mostly rodents) as intermediate hosts  guidelines when performing the literature search. Two researchers (T.Z. and B.L.) independently searched for relevant articles published in four English  and four Chinese  databases from their inception to August 22, 2020. The search terms were [(“echinococcosis” OR “echinococcoses” OR “echinococcus infection” OR “hydatidosis” OR “hydatidoses” OR “hydatid cyst” OR “hydatid disease” OR “echinococcus granulosus infection”) AND (“risk factor” OR “population at risk” OR “homo sapiens” OR “man” OR “human”) AND (“People's Republic of China” OR “Chinese” OR “China”)]. In addition, the references of reviews and meta-analyses were manually screened to identify additional potentially relevant studies.Studies were eligible if they met the following inclusion criteria: (1) the research was conducted with Chinese residents; (2) the diagnoses of AE and CE were based on a combination of serological and ultrasonic methods; and (3) the odds ratios (ORs) and their 95% confidence intervals (95% CIs) could be obtained directly or calculated from the study.Studies were excluded if (1) the publications were neither in Chinese nor in English; (2) the sample size was ≤ 30 ; (3) no Two authors (TZ and BL) independently assessed the quality of the studies. We employed NOS and AHRQ scores to assess the quality of cross-sectional studies and case-control studies. NOS scores range from 0 to 8; scores of 7–8, 4–6 and 0–3 indicate a study of high, medium and low quality, respectively , 14. AHRTwo authors (TZ and BL) independently extracted data and information from the studies including the first author, year of publication, region, type of echinococcosis (AE or CE), study design, sample size, number of positive cases, participant age in years, sex, participant race/ethnicity, whether the participants were herdsmen, ang/or raised dogs, the kind/number of animal hosts, participant hand washing status, and the OR value and its 95% CI, or the original data from which the OR could be calculated.2 statistic. An I2 < 25%, 25–50%, 50–75% and 75–100% represents no, moderate, large and extreme heterogeneity, respectively  and geographic distribution of the studies . Data were analyzed using the R (4.0.0) package meta , 21, andA total of 1,026 articles were originally identified; of these, 449 were excluded as duplicates. Thus, 577 studies were screened; of these, 26  were elin = 1), Qinghai Province (n = 4), Western China (n = 2), Yunnan Province (n = 1), Gansu Province (n = 3), Xinjiang Province (n = 5), Ningxia Province (n = 5), Sichuan Province (n = 3), and the Tibetan Plateau (n = 2). Three of the included studies reported only on AE-infected patients, seventeen reported only on CE-infected patients, and six studies reported on both AE-and CE-infected patients.Overall, the included studies covered 690,322 individuals , of which 38,358 had echinococcosis  according to the combined diagnosis based on ultrasound and serological methods. The included studies varied in location, including the Tibetan Autonomous Region  and 4.45% (95% CI: 2.53–7.71%), respectively.Seven risk factors for AE were indicated among the studies, and a meta-analysis was executed on nine cross-sectional studies , 42–45. p < 0.001), herdsman occupation , not washing hands before meals  and sex  .Four of these seven risk factors were statistically significant. They are listed according to the strength of the correlation as follows: ethnicity (Tibetan vs. Han) , nomadism  and drinking nonboiled water .All of these eleven risk factors were statistically significant. The top three, according to the strength of the correlation, were ethnicity (Tibetan vs. Han) ] The publication bias was assessed for all the risk factors included in this study see . Based op = 0.029) and feeding viscera to dogs . In the cross-sectional studies, seven risk factors were identified. The top three, according to the strength of the correlation, were ethnicity , not washing hands before meals  and herdsman occupation . The heterogeneity of all the CE risk factors decreased to varying degrees in the study design-specific subgroup analysis.The study design-specific subgroup analysis examined only risk factors for CE because all of the AE studies had a cross-sectional design. The results are shown in P = 0.006). The significant risk factors for CE were participant sex in Ningxia  and Qinghai  and herdsman occupation in Xinjiang . The In general, the studies were widely geographically distributed, including seven Chinese provinces. The provinces of Ningxia, Qinghai, and Xinjiang were included in the study region-specific subgroup analysis, and had three risk factors, two risk factors, and one risk factor, respectively. Only one risk factor for AE was identified  and 4.45% (95% CI: 2.53–7.71%), respectively, in the included individuals. However, the prevalence of AE was significantly higher than that previously calculated in China (0.96%) , demonstSex, ethnicity, not washing hands before meals and herdsman occupation were found to be common risk factors for AE and CE in this meta-analysis. Similar to previous studies , 49, we via the ingestion of food, soil, or water contaminated with the feces of infected mammals  that have inherent limits; for instance, observational studies are prone to selection bias and space bias . In addiIn summary, understanding the risk factors for echinococcosis provides a scientific basis to guide the formulation of prevention and control measures. Of the risk factors examined, for both AE and CE, the most important was ethnicity. Tibetans are at the highest risk of echinococcosis and thus must be closely monitored. The evidence for dog-related risk factors is also convincing, albeit at a lower level than that of ethnicity. Preventative measures of echinococcosis in humans should aim at raising the awareness of the disease in target groups and dog management. A series of national control measures, including regular dog deworming, public health education and community screening, should be implemented.The original contributions presented in the study are included in the article/TZ and SL: study design and drafting of the manuscript. TZ, BL, and YL: data collection, analysis, and interpretation. All authors approval of the final version for publication.The study was supported by the National Natural Science Foundation of China (81860606) and the Natural Science Foundation of Qinghai Province (2019-ZJ-906).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.


A conflicting body of evidence suggests localized periodontal inflammation spreads systemically during pregnancy inducing adverse pregnancy outcomes. This systematic review and meta-analysis aim to specifically evaluate the relationship between periodontitis and preeclampsia.Electronic searches were carried out in Medline, Pubmed, Embase, Lilacs, Cochrane Controlled Clinical Trial Register, CINAHL, ClinicalTrials.gov, and Google Scholar with no restrictions on the year of publication. We identified and selected observational case–control and cohort studies that analyzed the association between periodontal disease and preeclampsia. This meta-analysis was conducted following the PRISMA checklist and MOOSE checklist. Pooled odds ratios, mean difference, and 95% confidence intervals were calculated using the random effect model. Heterogeneity was tested with Cochran’s Q statistic.Thirty studies including six cohort- and twenty-four case–control studies were selected. Periodontitis was significantly associated with increased risk for preeclampsia , especially in a subgroup analysis including cohort studies . The association was even stronger in a subgroup analysis with lower-middle-income countries .Periodontitis appears as a significant risk factor for preeclampsia, which might be even more pronounced in lower-middle-income countries. Future studies to investigate if maternal amelioration of periodontitis prevents preeclampsia might be warranted. The most recent systematic review and meta-analysis on the relationship between periodontal disease and preeclampsia were published in 2013 and could only detect the statistical differences in a subgroup of case-control studies. For that reason, we conducted this study to re-evaluate this association, especially with regard to cohort study design. Furthermore, we take socioeconomic factors into consideration. This systematic review and meta-analysis not only empower the positive association between periodontitis and preeclampsia during pregnancy by the larger sample size and the data synthesis of cohort studies but also point out the considerable difference in lower-middle-income countries.Preeclampsia is the onset of pregnancy-related hypertensive disorder and proteinuria arising most commonly after 20 weeks of gestation, which could lead to eclampsia and induce maternal and perinatal morbidity, and mortality. The prevalence of preeclampsia is between 2 to 8% of all pregnancies worldwide Duley, . PreeclaContributing to USD 6.4 billion short-term estimated costs for preeclamptic pregnancies in US healthcare system, USD 1.03 billion were spent on maternal healthcare and USD 1.15 billion were expended for infants born to these women while the remaining expenses were for peripartum and postpartum care  guidelines with the checklist of 27 items to conduct our study . The reference lists of relevant articles were also scanned for appropriate studies. No language restrictions were adopted in either the search or study selection. No search for unpublished literature was carried out. Authors were contacted for translation and information.Two independent reviewers (LD and HN) reviewed the titles, abstracts and methods of retrieved results to assess for the eligibility criteria. When there was a disagreement in a selection process between reviewers, consensus with the third reviewer (QA) was obtained.Low income: $995 or less;Lower-middle-income: $996–$3945;Upper-middle-income: $3946–$12,195;High income: $12,196 or more.Data extraction was carried out using a standardized extraction form, collecting information on the first author’s name, publication year, study design, number of cases, number of controls, total sample size, country, national income group, mean age, the risk of estimates or data used to calculate the risk estimates, CIs or data used to generate CI. According to the World Bank classification of countries which is based on Gross National Income per capita, groups of national income per year are  , total number of participants per group, overall Odds Ratio and Mean difference.Data were imported in a statistical software . Pooled Odds Ratios, mean difference, and 95% Confidence Intervals were calculated for the association between periodontitis and preeclampsia using a random effects model. The pooled effect was considered significant if p-value was less than 0.05. Forest plots for primary analysis and subgroup analysis show the raw data, Odds Ratio and CIs, Means and SDs for the chosen effect, heterogeneity statistic , definition of periodontitis (defined by pocket depth (PD) and/or clinical attachment loss ), mean CAL, mean PD, national income (high-income or middle-income or low-income countries).2 statistic, which represents the percentage of the total variability across studies which is due to heterogeneity. I2 values of 25, 50 and 75% corresponded to low, moderate and high degrees of heterogeneity respectively  revealing a significant variation among studies.The results of the meta-analysis showed that periodontitis was associated with increased risk for preeclampsia  but high for case–control .According to the study type, the results revealed the increased risk of preeclampsia in periodontitis patients in the cohort studies   and CAL alone , while significantly high heterogeneity was found in the subgroup which periodontitis was defined by CAL and PD  , CAL and PD , CAL alone . Heterogeneity was moderate for the subgroups in which periodontitis was defined by PD alone  . Likewise, the preeclamptic group had a statistically higher mean PD compared to healthy group . The heterogeneity was 98% in both subgroup analysis, (I1) Figs. .Fig. 10FThe funnel plot for the association between periodontitis and preeclampsia revealed the symmetry Fig. . No publThe aim of this meta-analysis was to re-evaluate the potential association between preeclampsia and periodontitis. The results confirm that periodontitis is a risk factor for preeclampsia, which was similar to the finding of a meta-analysis in 2013 by Sgolastra et al.  was substantially lower than in a study by Sgolastra et al. , again strengthening the reliability of our results  or one site with pocket depth ≥ 5 mm . Moreover, the subgroup analysis with lower-middle-income countries, which generated an Odds ratio of 6.70, indicated the considerable Socioeconomic inequalities in access to oral health care accounted for 60% in lower-middle-income countries (Hosseinpoor et al., We used Newcastle Ottawa Scale to evaluate the risk of bias and found twenty studies with moderate risk of bias and ten remaining studies with a high risk of bias. Exposure bias in this study was due to the non-response rate which was not described clearly in these studies. No publication bias was detected. Sensitivity analysis resulted in no change to the finding of the study.The strength of this systematic review and meta-analysis includes the large sample size of 9650 subjects. Six cohort studies comprising 2840 subjects were analyzed and revealed the statistically significant difference. Because systematic reviews of prospective cohort studies generate more reliable evidence, our study provided an updated systematic review and meta-analysis and confirmed the association between periodontitis and preeclampsia (Hillier et al., This meta-analysis confirms previous findings of an association between periodontitis and preeclampsia. However, this study includes fifteen more recent publications, which resulted in a larger effect size of the association, specifically, for lower-middle-income countries in comparison to high and upper-middle-income countries. Our results warrant future studies to investigate the mechanisms of this association and whether targeted interventions to prevent or treat periodontitis preconception or during pregnancy can lead to better pregnancy outcomes.


Atopic diseases and atrial fibrillation (AF) seem to share an underlying inflammatory pathology. To date, some population-based studies have explored the relationship between the two. We aimed to conduct a meta-analysis to examine the role of atopic condition in AF risk.2 and Cochran Q test were used to evaluate the heterogeneity.All relevant observational studies in PubMed and EMBASE databases up to November 2021 were searched. In RevMan 5.3, we used random-effects or fixed-effects models to pool the effect sizes of hazard ratio (HR), odds ratio (OR) and their corresponding 95% confidence intervals (95% CI). In addition, In = 4; 95%CI, 1.25–1.58), 1.12  and 1.06 , respectively.A total of 2488 records were retrieved. After screening according to the predetermined criteria, 6 cohort studies and 2 case-control studies were included in this meta-analysis. Herein, the meta-analysis of 6 cohort studies suggested that atopic diseases potentially increased the AF risk with the pooled HR of 1.26 , while the pooled effect size  of 2 case-control studies was not statistically significant. Based on the types of atopic diseases, further subgroup analyses of 6 cohort studies revealed that asthma, allergic rhinitis, and atopic dermatitis all potentially increased the risk of subsequent AF with the pooled HR of 1.41 (This meta-analysis demonstrated that patients with atopic diseases have a higher risk of developing AF, particularly those with asthma. Atrial fibrillation (AF) is the most common type of cardiac arrhythmia in adults, with an increasing trend in both morbidity and mortality . The recAtopic diseases are primarily composed of asthma, allergic rhinitis, and atopic dermatitis, and may coexist in the same individuals. Specifically, allergic rhinitis mainly involves the upper airway and is characterized by nasal congestion, rhinorrhea, sneezing, and nasal itching . Asthma Emerging evidence suggested that inflammation and immune response are involved in the initiation and maintenance of AF . MultiplThe statement of Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) was followed to perform the meta-analysis . BesidesTwo investigators  independently retrieved relevant articles in PubMed and EMBASE from inception to the search end . The search terms were mainly involved with “atrial fibrillation” and “atopic diseases.” We did not restrict language, region of study, or sample size for publications retrieved. The details of our search strategy are summarized in Inclusion criteria: (1) observational (cohort or case-control) studies; (2) cohort studies must include patients with atopic diseases and non-atopic individuals. And, the relative risk (RR) or hazard risk ratio (HR) with 95% confidence intervals (95% CI) of incident AF comparing between patients with atopic diseases and individuals without atopic diseases must be reported. (3) case-control studies must consist of patients with AF and controls. Besides, the studies also need to trace their allergic history and provide odds ratio (OR) with 95%CI of the association between atopic diseases and AF. Exclusion criteria: (1) basic experimental researches; (2) lacking or not available data; (3) case reports, reviews, letters, replies, or comments; (4) duplicate publications of the same database.For all qualified studies, a standardized collection list was used to extract the following details: first author’s name, publication year, country of the study, study design, the age of participants, sample size, the approaches of disease definition, variables adjusted in multivariate analysis.Next, the Newcastle-Ottawa Scale (NOS) was used to evaluate the quality of all included studies. For any controversial contents, the three researchers  discussed and made decisions together. The details of scoring were outlined in 2, respectively. In detail, the value of I2 was divided into 4 grades, namely high heterogeneity (>75%), moderate heterogeneity (51–75%), low heterogeneity (26–50%), and insignificant heterogeneity (<25%) (I2 > 50%), a random effect model was used to pool the effect size. What’s more, we further used sensitivity analysis to find the sources of high heterogeneity.The effect sizes of HR, OR, and their corresponding 95%CI were extracted as the primary analysis statistics. In addition, based on study design and the types of atopic diseases, we performed subgroup analyses to clarify the association between atopic diseases and the risk of AF. For the heterogeneity of included studies, we assessed it qualitatively and quantitatively using Cochrane Q test and Iy (<25%) . During 1 and STATA 15.0 . Two-sided p < 0.05 was viewed as statistically significant.In this study, data were analyzed using the software of RevMan 5.3A total of 2488 records were retrieved from the databases of PubMed and EMBASE. After eliminating duplicates, the 2353 records were reviewed with titles and abstracts. After initial screening, 2312 articles were excluded because they were inconsistent with the scheduled criteria. Then, we conducted a comprehensive review of the full text of 41 articles and eliminated 33 articles because of insufficient results, or repeated publication of the same database. Eventually, a total of 8 studies (6 cohort studies and 2 case-control studies) , 18–23 wThe included six cohort studies consisted of 1,452,112 patients with atopic diseases and 7,373,745 comparators. Three studies were conducted in Europe, 2 in Asia, and 1 in North America. Regarding specific types of atopic diseases, 4 studies investigated the association between asthma and AF, 3 studies examined the relationship between atopic dermatitis and AF, and another study involved allergic rhinitis and AF. All 6 studies extracted the effect size of HR, with 5 providing adjusted HR and 1 providing crude HR.Among the 2 case-control studies, one from Asia (China), and the other from North America (United States). In total, they traced the history of asthma in 8,869 patients with AF and 11,505 controls, and used the adjusted OR (aOR) to assess the relationship between the two.Both cohort and case-control studies mainly focused on adult patients, and the definitions of AF and atopic diseases  were based on matching International Code Designator, documentation of medication, self-reported or physician recorded diagnosis of disease. Moreover, based on the four characteristics: daytime symptoms (2 times per week or less or more than 2 times per week), night awakenings (none or any), the need for reliever medication (2 times per week or less or more than 2 times per week), and limitation of activities (none or any), Cepelis et al.  classifiI2 = 95%, p < 0.001) ; nevertheless, the high heterogeneity was observed (< 0.001) .2 of 78% , and the heterogeneity was also high with I2 of 78% .I2 = 75%) (I2 = 0%)  . For atoI2 = 0%) . Only onI2 = 0%) .I2 = 0%) and 1.34 , respectively .Given the limited number of included studies, we assessed publication bias in the subgroups of cohort studies and asthma cohort studies. There was no obvious publication bias in either subgroup , especially those in persistent or uncontrolled condition. Allergic rhinitis and atopic dermatitis were second to asthma in their association with the risk of new-onset AF .Currently, the potential mechanism that could explain the association between atopic diseases and AF may be chronic systemic inflammation. As we all know, after exposure to environmental allergens, epithelial-derived alarmins  could activate T helper type 2 (Th2) cells and group 2 innate lymphoid cells (ILC2s). Then these immune cells would release a series of cytokines, such as IL-4, IL-5, and IL-13, which promote the production of immunoglobulin E (IgE), eosinophilic inflammation . In addiSecondly, mast cells, a key regulator of allergy and inflammation, are also abundant in the human heart . When alIn this meta-analysis, asthmatics were identified as having a higher risk of developing AF. Due to the different definitions of asthma subtypes, we did not perform relevant subgroup analyses. But it is worth noting that Cepelis et al.  found a The relationship between atopic dermatitis and cardiovascular disease is controversial . Our metThere are some limitations should be considered when the results of our meta-analysis were interpreted. First, the small number of studies available for inclusion limited the possibility of conducting more subgroup analyses to some extent. Second, the definition of allergic diseases and AF lacks preciseness and specificity. For instance, all the included studies did not distinguish between paroxysmal and persistent AF, so we cannot make a further clinical interpretation. Third, the medication of patients with atopic diseases is a factor that must be considered for their secondary diseases. However, only 2 studies (one cohort study and one case-control study) performed subgroup analyses of the association between medications (β-agonist and corticosteroid) for asthma control and AF risk and demonstrated the positive relationship between the two. In the future, it is necessary to obtain relevant clinical information to clarify the causes of AF in patients with atopic diseases. Besides, the study of Yang et al. 2011  that hadIn summary, three atopic diseases  were noted to increase the risk of AF. In the future, more high-quality cohort and case-control studies should be conducted to confirm the association between atopic diseases and AF, with particular attention to whether the increased risk of different types of AF is related to the type, severity, and treatment of atopic diseases.The original contributions presented in this study are included in the article/RZ, JW, JZ, and LD proposed and designed the study. RZ, JW, and ZL contributed to the data collection, statistical analysis, and quality assessment. RZ and JW interpreted the data and wrote the original manuscript of the study. ZL, JZ, ZW, and CX conducted major revisions. All authors critically revised important intellectual content of the study and approved the final version for publication.


TPH1 polymorphisms with the risk of suicide behavior (SB).We aimed to examine the association of Systematic review and meta-analysis.TPH1 gene polymorphisms and SB was evaluated using inherence models by odds ratio (OR) and 95% confidence interval (CI). Subgroup analyses, heterogeneity analyses, and publication bias were also tested in this meta-analysis.All relevant studies that evaluated the association between the A218C (rs1800532), A779C (rs1799913) and A6526G (rs4537731) polymorphisms and the susceptibility to SB published up to September 2021 were identified through a comprehensive systematic search in PubMed, Scopus, EBSCO and Science Direct electronic databases. The association between TPH1 A218C revealed an increased risk of SB in the dominant model . We also observed a positive association in the allelic , homozygous , heterozygous  and dominant  inherence models with the suicide attempt. Additionally, in the heterozygous  and dominant  inherence models we detected an association with completed suicide. Based on ethnicity, an association of SB in the European population also was observed . However, for both A779C and A6526G polymorphisms we did not find evidence of an association with SB.The meta-analysis for TPH1 gene could be a possible risk factor of SB. Future large-scale studies are required to analyze the molecular mechanisms by which affect the susceptibility of developing suicide behavior.This meta-analysis suggests that the A218C polymorphism of  Human behavior is governed by a complex interplay of numerous biological, genetic, psychological, cultural, social and family determinants. In particular, suicide behavior (SB) ranges from suicide ideation to suicide attempts and completed suicide and constitutes a multifactorial public health issue. The understanding of the genetic system that causes vulnerability to develop SB is largely incomplete, as its etiology is complex and diverse; however, epidemiological studies show that suicide behavior is partly heritable and polygenic , 2.Serotonin, also known as 5-hydroxytryptamine (5-HT), is produced by generating 5-hydroxytryptophan out of tryptophan and the downstream decarboxylation using aromatic amino acid decarboxylase . TryptopThe serotonergic system is implicated in the etiology and pathogenesis of several psychiatric disorders. Serotonergic pathways associated with mood disorders and SB include the inflammatory processes related to serotonin processing . The neuTPH1 gene has been extensively studied as a candidate for suicide behavior due to its role in serotonergic neurotransmission  reporting guidelines.The search strategy follows the PRISMA  AND (“TPH1” OR “tryptophan hydroxylase 1”). We searched for articles were searched in PubMed, Scopus, EBSCO and Science Direct databases, dated up to September 2021. To find other potential studies, we examined the references of the eligible studies. The process of the study selection in the preset work is detailed in a flow diagram; TPH1 A218C, A779C and A6526G polymorphisms and suicide risk, 2- cases-controls designs studies, 3- individuals with a suicidal behavior which identified by a specialist; we included all the suicidal behavior spectrum  addressed in the published studies, 4- sufficient data available to obtain genotypic frequencies to calculate odds ratio (OR) and 95% confidence interval (95%CI), 5- articles published in English. These data were not always available for all studies, but when needed, we contacted the authors to clarify the information not included in the papers. Reviews, meta-analysis, duplicates, case reports, book chapters, and animal studies were all excluded.The criteria applied for eligible studies were: 1- studies that evaluated the association between The relevant data from each study were extracted by to reviewers (Hernandez-Díaz and Castillo-Avila) using standardized and structured forms. The following data were extracted: first author, year of publication, country, main outcome, criteria diagnostic, source of DNA, genotyping method, ethnicity, gender, sample size, genotype distribution for the cases and controls. Once encountering discrepancies, other authors re-checked the original articles until an agreement was achieved by all of them.The Newcastle-Ottawa Scale (NOS) was used to examine the quality of each study. Quality scores range from 0 to 9, and higher scores mean better quality of the study. Studies scoring six o higher were considered as high-quality studies. The qualities of the includes studies were evaluated by the same two investigators. Disagreements were resolved through discussion by a third investigator .TPH1 gene polymorphism and SB risk was determined by calculating OR and 95%CI. To measure this association, we evaluated the effect by five genetic models: allelic (A vs. C), dominant (AA + AC vs. CC), recessive (AA vs. AC + CC), heterozygous (AC vs. CC) and homozygous (AA vs. CC).The relationship between P < 0.05 was considered statistically significant). Cochrane Q-test and I2 test was conducted to assess the between studies heterogeneity. Significant heterogeneity was defined with a P-value < 0.05 and I2 ≥ 50% and a random effect model (DerSimonian–Laird) was used in this case. Otherwise, a fixed effect model (Mantel–Haenszel) was applied. Subgroup analysis was used to further identify the factors influencing heterogeneity. Publication bias was assessed by funnel diagram that produces a diagram according to its OR and the standard error of each study, moreover, Egger's test was calculated . The data were analyzed using Comprehensive Meta-Analysis (CMA) software v.2.Z-test was used to determine the significance of the OR , suicide ideation (n = 1), and completed suicide (n = 8). The age of the individuals studied ranged from 17.4 to 65 years old. There were statistically significant deviations from the HWE in the control groups (n = 3); therefore, these studies were excluded from the meta-analysis.There were 6,214 cases and 12,184 controls in these 36 studies, 16 studies were conducted in European populations, 13 in Asian populations, 5 in American populations, 1 in Turkish and 1 in mixed populations . In terms of the SB type, some studies evaluated suicide attempt   assessed the relationship between behavior , 45–47. = 19.90) .P = 0.000, Q test = 0.233, I2 = 19.72), homozygous , heterozygous  and dominant  inherence models with the suicide attempt; while a decreased risk of completed suicide was observed in the heterozygous model  and dominant model  . However, no significant associations were found in Asian populations.Subgroup analyses were then performed based on ethnicity and diagnostic in order to investigate sources of heterogeneity . We obse= 14.38) . In the TPH1 A779C polymorphism and the risk of suicide behavior  assessed relationship between behavior , 46, 47. SB risk . Either TPH1 gene A6526G polymorphism was not significantly associated with suicide behavior under any genetic model.Three case–control studies with 516 patients and 441 controls were included in the present meta-analysis , 26, 35.P-values demonstrated nonsignificant values ranging from 0.15 to 0.99.We estimated potential publication bias in this meta-analysis with funnel plots and Egger's test. Funnel plots were found to be overall symmetrical , 3B and TPH1 gene and the risk of suicide behavior. Considerable evidence has shown that the TPH1 gene is a possible candidate involved in the etiology of suicide. Although one meta-analyses  metabolism has two large pathways: the methoxyindole pathway and the kynurenine pathway. Regarding the available TRYs in the body, approximately 1~5% are synthesized as serotonin through the methoxyindole pathway and 95~99% of TRYs are metabolized through the kynurenine pathway and form tryptophan catabolites (TRYCATs) , 51. TheMoreover, this polymorphism could also influence individual differences in anger-related personality traits. Individuals carrying the AA genotype have a reduced capacity of controlling anger expression when they experience this emotion. Conversely, the alternative genotypes (AC and CC) can be considered as “protective” against the tendency to lose control when experiencing anger . TherefoThe observed differences in susceptibility, between suicide attempt and completed suicide, are likely due to the overall genetic background that modifies the SB prone risk factors. Moreover, this discrepancy in SB risk may be explained by daily lifestyle, geographic climate, dietary habits, ethnic diversity and so on. However, these results should be illustrated prudently and need further confirmation by more trials.TPH1 and other genes on SB susceptibility, such as TPH2. Therefore, it is necessary to systematically screen for functional variants within the TPH1 gene and other related genes in the SB pathogenesis.Second, A779C and A6526G polymorphisms were not associated with the risk of suicide behavior. Still, we cannot ignore the interaction between TPH1 polymorphism and the risk of SB and analyzed the role A6526G polymorphism on suicide behavior for the first time. Methodological issues were well explored  in the present work. Last but not least, the present work was carried out with five genetic models and sub-analyses by population and diagnostic subgroups were performed. Suicide attempt and completed suicide were not analyzed in previous meta-analyses.In González-Castro  meta-anaTPH1 and suicide ideation. Moreover, the inclusion of the studies that dealt with only ideation may have increased the heterogeneity of the analyzed data.The current meta-analysis also has some limitations. First, only articles published in English-language journals were included. Second, inter-gene and gene–environment interactions might also influence the accuracy of our outcomes. A lack of the original data restricted further evaluations of the potential inter-gene and gene–environment interactions. Related to, we recognized as a limitation that considerable percent of the studies were performed in some specific population (e.g. Croatian or German), this possible overlap of clinic center and sample population could introduce bias in the outcome. Therefore, the findings should be taken with caution. Finally, in the suicide ideation group, the number of relevant original documents was limited, there was not enough data to identify the relationship between TPH1 polymorphisms and suicide behavior. This meta-analysis revealed a significant association between the A218C polymorphism of TPH1 gene and SB. However, neither in overall population nor in subgroup analysis was found a significant association between A779C and A6526G polymorphisms and SB susceptibility. Therefore, the A218C polymorphism could be considered as one possible risk factor of SB. Further studies that investigate the relative contribution of TPH1 polymorphisms and the mechanisms by which they affect the pathogenesis of SB are required to corroborate these findings.The current meta-analysis gives a comprehensive analysis of the available information for the association between the The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation.Conceptualization, investigation, methodology, validation, visualization, roles/writing—original draft, and writing—review and editing: YH-D, TG-C, CT-Z, RC-A, ML-N, MR-M, and AG-M. Data curation: YH-D, TG-C, CT-Z, and RC-A. Formal analysis and resources: TG-C, CT-Z, ML-N, and AG-M. Funding acquisition: YH-D, TG-C, CT-Z, and ML-N. Project administration: YH-D, TG-C, and CT-Z. Software: YH-D, TG-C, and CT-Z. Supervision: CT-Z, RC-A, ML-N, MR-M, and AG-M. 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.