In a survey and three experiments , we examined if vaccination requirements are likely to backfire, as commonly feared. We investigated if relative to encouraging free choice in vaccination, requiring a vaccine weakens or strengthens vaccination intentions, both in general and among individuals with a predisposition to experience psychological reactance. In the four studies, compared to free choice, requirements strengthened vaccination intentions across racial and ethnic groups, across studies, and across levels of trait psychological reactance. The results consistently suggest that fears of a backlash against vaccine mandates may be unfounded and that requirements will promote COVID-19 vaccine uptake in the United States. Will requiring a behavior create resistance to it or successfully promote it? What are the psychological implications of requiring a behavior like vaccination? In particular, should we require vaccination for COVID-19 , or should we simply make a recommendation for employees, students, or customers to vaccinate if they so choose?3. According to this literature, people may simply conform to what is required because they fear social sanctions or view themselves as ethical individuals who follow norms4. They may also perceive the requirement as beneficial for them and society at large .Two important streams of research make contrasting assumptions about the impact of requiring people to perform a behavior, such as vaccination. In policy and law scholarship, requirements are expected to guide behavior12. As implied by the theory of psychological reactance16, people may resist a requirement they perceive as curtailing their freedom2. In a review of the literature on COVID-19 vaccination hesitancy, Lin et al.17 warned against mandates arguing that prior resistance to mask mandates suggests that vaccine mandates may also elicit reluctance. Specifically, they wrote: “Framing vaccination as a smart, purposeful personal decision, emphasizing individual’s autonomy could yield greater results.” They went on to state that mandates are likely to elicit resistance “even among originally receptive groups”17. Other scholars have cautioned that “considerable part of the target group could be annoyed when the act comes into force, consequently showing reactance and opting out of other vaccines that are still voluntary,” concluding that a well-meant mandate could do more harm than good16.In contrast to the above literature, research in psychology and other behavioral sciences often views requirements with skepticism12. A negative effect was only found in an experiment using a computer vaccination game that forced players to vaccinate against their will, which in turn led them to vaccinate less at a later point of the computer game12. Thus, we conducted four studies that provide this evidence.The experimental evidence about the impact of vaccination requirements, however, has been limited. A study on mandating a hypothetical children vaccine in an experimental context found no impact on the vaccination intentions of a sample of American parentsOur first three studies were within-subjects experimental to ensure that intentions to vaccinate varied as a function of requirement and freedom of choice rather than as a result of idiosyncratic characteristics of the participants. The first included a survey of White American, Black American, and Hispanic American respondents that examined the degree to which requiring the COVID-19 vaccine to work, travel, or go to school would increase or decrease intentions to vaccinate. This study was followed by three experiments, the final one preregistered, comparing different scenarios about a vaccination decision in the workplace. Respondents considered the situation in which either a new employer (Studies 2 and 3) or their current employer (Study 4) tells them that (a) “they require that you get vaccinated against a new disease” (requirement condition), (b) “they prefer that you get vaccinated against a new disease but that vaccination is up to you” (freedom of choice condition), and (c) “getting vaccinated against a new disease would give you more freedom and flexibility” .In each experiment, we examined the impact of the requirement manipulation on vaccination intentions. In Study 4, we also examined the impact of the requirement manipulation on perceived vaccination benefit , perceived fairness , perceived obligation , and perceived vaccination norms . In addition, we examined whether the requirement affected participants with different levels of psychological reactance (Studies 3 and 4), which is the principal reason why requirements could backfire. We also studied whether norms moderated the impact of the requirement (Study 3).https://osf.io/y2rjn/?view_only=1f313bc39b234ed4be0d30c44d8ff126. Each study and its results are summarized below. For additionall information, such as materials and verbatim items, statistical power calculations, and demographic characteristics, please see the Methods section at the end of the manuscript and a summary of the studies in Table Norms and psychological reactance were important to study because they might be preconditions for negative effects of vaccination requirements. Individuals with higher levels of trait psychological reactance may resent the requirement and thus form weaker intentions to vaccinate, particularly if vaccination norms are weak. The data and codes from our studies are available in N = 108), Black (N = 105), or Hispanic (N = 96) respondents living in the US recruited from the platform Prolific (www.prolific.com) in January 2021. The survey included two questions: “Will you get the COVID-19 vaccine if it is required to work, travel, or go to school?” (requirement question) and “If you could get the COVID-19 vaccine for free today, would you want to be vaccinated today?” (control question), both answered as either “Yes” or “No.”Study 1 consisted of a survey of 299 participants who were selected to include Non-Hispanic White (n = 10), leaving us with 105 white-only, 95 black-only, and 89 hispanic-only participants. The results for each racial/ethnic group were similar to the results for the population as a whole. Eighty six percent of white-only respondents reported intending to vaccinate in response to the required vaccination question but only 69 percent of them reported intending to vaccinate in response to the control vaccination question. Ninety four percent of Hispanic-only respondents reported intending to vaccinate in response to the required vaccination question but only 80 percent of them reported intending to vaccinate in response to the control vaccination question. Eighty percent of Black-only respondents reported intending to vaccinate in response to the required vaccination question but only 56 percent of them reported intending to vaccinate in response to the control vaccination question.Analyses of these questions using a two-sided chi-square test appear in Table https://www.mturk.com/) and entailed a 3-cell, within-subjects experimental design in which participants rotated through each condition and conditions were presented in random order. Participants were told that they have a new job and the company tells them that (a) “they require that you get vaccinated against a new disease” (requirement condition), (b) “they prefer that you get vaccinated against a new disease but that vaccination is up to you” (freedom of choice condition), and (c) “getting vaccinated against a new disease would give you more freedom and flexibility” (condition that emphasizes freedom as a vaccination outcome and thus controls for freedom of choice). The dependent measure was the intention to vaccinate, which was measured on a scale from 1 (extremely unlikely) to 5 (extremely likely).Study 2 was conducted in March 2021 with 359 participants recruited from Mechanical Turk  = 6.98, p = .001. Specifically, intentions were significantly stronger in the required vaccination condition  than either the freedom condition , t(716) = 3.63, p = .0009, Mdiff = 0.18, SDdiff = 0.05, 95% CI, d = 0.14, or the control for choice freedom condition , t(716) = 2.59, p = .026, Mdiff = 0.13, SDdiff = 0.05, 95% CI , d = 0.10. As a reminder, intentions were measured on a 1 to 5 scale, where 1 = extremely unlikely, 2 = somewhat unlikely, 3 = neither likely nor unlikely, 4 = somewhat likely, and 5 = extremely likely. Therefore, the requirement moved participants closer to the likely point of the scale.Analyses of variance of intentions as a function of experimental condition showed that vaccination intentions differed as a function of our experimental manipulation; omnibus https://www.mturk.com/) and entailed a 3  × 2 (norms: 70% of employees are vaccinated versus 30% of employees are vaccinated) within-subjects factorial design. This design ensured that all participants rotated through all conditions and the order of the conditions was random.Study 3 was conducted in March 2021 and examined the boundaries of our effect. This study involved 357 participants recruited from Mechanical Turk  positively correlate with intention to vaccinate against HPV (Human Papilloma Virus)F = 10.57, p < .0001. That is, intentions were stronger in the requirement condition than in the freedom of choice condition, t(712) = 4.42, p < .0001, Mdiff = 0.15, SDdiff = 0.03, 95% CI, d = 0.17, and the control for choice freedom condition, t(712) = 3.31, p = .003, Mdiff = 0.11, SDdiff = 0.03, 95% CI, d = 0.12. Moreover, the manipulation of norms also had an impact, such that participants who were told that 70 percent of the employees were vaccinated formed stronger intentions to vaccinate than did those who were told that only 30 percent of the employees were vaccinated; F  = 33.55, p < .0001, Mdiff = 0.19, SDdiff = 0.03, 95% CI , d = 0.31. In addition, however, intentions to vaccinate were a function of the interaction between the vaccine requirement manipulation, norms, and psychological reactance, F = 3.19, p = .04. There was also a significant main effect of psychological reactance  but no significant interactions between requirement condition and reactance , or between requirement condition and norm .The results from Study 3 appear in Table F for the simple effect of the requirement manipulation = 7.78, p = .0005, regardless of participants’ psychological reactance, F for the interaction between requirement and psychological reactance = 0.33, p = .72. Specifically, in these more positive norm conditions, the requirement condition produced stronger intentions than the freedom condition. t(712) = 3.85, p = .0004, Mdiff = 0.16, SDdiff = 0.04, 95% CI, d = 0.14, and the control for choice freedom condition, t(712) = 2.70, p = .02, Mdiff = 0.11, SDdiff = 0.04, 95% CI, d = 0.10.To further explore the interaction between the requirement manipulation, norms, and psychological reactance, we examined the interaction between the vaccine requirement and psychological reactance for the lower and higher norm conditions separately. When 70% of the population was vaccinated, requiring the vaccine had an advantage over not requiring it, F for the simple effect of the requirement manipulation = 5.02, p = .007, but this effect differed for participants with lower and higher levels of psychological reactance, F for the interaction between the requirement manipulation and psychological reactance = 3.02, p = .049. Specifically, when only 30% of the population supported vaccination and participants had lower reactance (− 1SD of the mean), requiring the vaccine had no effect on intentions (requirement vs. control for choice freedom: t(1775) = 1.67, p = .10; requirement vs. freedom: t(1775) = .60, p = .55). When only 30% of the population supported vaccination and participants had higher reactance (+ 1SD of the mean), the requirement condition produced stronger intentions than the freedom condition t(1775) = 3.45, p = .0006, Mdiff = 0.23, SDdiff = 0.07, 95% CI, d = 0.16 but was similar to the control for choice freedom condition, t(1775) = 1.61, p = .11. Overall, we found no evidence that requiring the vaccine undermines vaccination intentions, even when examining its effects across different levels of vaccination norms and psychological reactance.Now, when only 30% of the population supported vaccination, requiring the vaccine also had an advantage over not requiring it, https://aspredicted.org/blind.php?x=6sb83y) administered to a nationally representative sample of United States adults in April 2021. This experiment involved 606 participants from Qualtrics panels  recruited with quotas for gender, education, and race/ethnicity, who provided complete data and passed the quality checks instituted by Qualtrics. The sample demographics appear in Table Study 4 was a preregistered experiment  perceived benefits of vaccination, (b) perceived norms for vaccination, (b) perceived fairness of vaccination, and (d) perceived obligation to vaccinate. As in Study Intentions as well as the four possible mediators of perceived benefits, perceived norms, perceived fairness, and perceived obligation were examined using analysis of co-variance. Because we recruited a nationally representative sample with potentially large demographic and behavioral variability, we included demographic covariates  as well as two indicator variables representing receiving the flu shot during the 2020–2021 season and receiving at least one shot of a COVID-19 vaccine. We also included psychological reactance and the interaction between our manipulation and psychological reactance, which was a preregistered exploratory analysis.M = 3.59, SE = 0.07) than describing vaccination as a free choice for participants to make , F = 6.24, p = .01, Mdiff = 0.23, SDdiff = 0.09, 95% CI, d = 0.10. This effect was present regardless of participants’ level of psychological reactance, F for the interaction between requirement and psychological reactance = 0.04, p = .85.The results from this study appear in Table F = 0.07 for perceived benefits, F = 0.01 for perceived norms, F = 2.37 for perceived obligation, and F = 0.07 for perceived fairness; p > .12 in all cases. The effects of our measure of psychological reactance were F = 1.74, ns, for perceived benefits, F = 2.92, ns for perceived norms, F = 4.25 p = .04 for perceived obligation, and F = 4.77 p = .03 for perceived fairness. The interactions between condition and psychological reactance were significant for perceived benefits, F = 6.98, p = .01 and perceived obligation F = 4.46, p = .035, but not for either perceived fairness, F = 2.15, ns, or perceived norm, F  = 1.75, ns.We also analyzed the effects of the experimental manipulation on perceived benefits, perceived norms, perceived fairness, and perceived obligation. The main effects of our manipulation were not significant in any case: F = 6.98, p = .01, for perceived benefits and F = 4.46, p = .035 for perceived obligation. A decomposition of these interactions showed that when participants had lower (1 SD below the mean) reactance, they perceived more benefits in the requirement condition  than the freedom of choice condition  = 2.08, p = .037, d = 0.09. However, their perceived obligation did not differ across the requirement  and freedom of choice conditions ; t(597) = 0.42, p = .68. In contrast, when participants were higher (1 SD above the mean) in reactance, they perceived similar benefits in the requirement condition  and the freedom of choice condition ; t(597) = 1.65, p = .10. However, their perceived obligation was lesser in the requirement condition  than the freedom of choice one ; t(597) =  − 2.56, p = .011, Mdiff = 0.31, SDdiff = 0.12, 95% CI, d = 0.10. Mediational analyses to interpret these differences are presented in In sum, requiring the vaccine continued to strengthen vaccination intentions and did so independently of psychological reactance. Also, requiring the vaccine did not directly influence perceived benefits, perceived norms, perceived fairness, or perceived obligation. However, we were also interested in the possibility that the mediational pathways from the requirement to intentions might differ across levels of psychological reactance, a possibility supported by the finding that the requirement condition interacted with participants’ level of reactance for perceived benefits and perceived obligation, which, as mentioned above, were Our surveys and experiments consistently suggested that making vaccination a requirement for work, attending school, or travelling is likely to be successful in promoting vaccination, more so than either giving people freedom to choose or reminding them of the freedom the vaccine might confer. Also, we found no evidence that requirements elicited reactance in a way that undermined the likely efficacy of a vaccine mandate. Rather, the behavioral intentions of participants who scored higher in trait psychological reactance were either more positively influenced by the requirement (see Study 3) or as influenced as their lower reactance counterparts (see Study 4). Study 4 also showed that participants who were higher in psychological reactance felt less obligated to vaccinate when the vaccine was required than when it was not. This result is intriguing and suggests that psychological reactance might lead people to feel less morally obligated to follow a mandate but still be motivated to vaccinate. All in all, our results suggest that fears about a mandate backfiring may be unfounded and support vaccine mandates as a strategy for achieving herd immunity for COVID-19 in the United States.N = 108), Black (N = 105), and Hispanic (N = 96) respondents. A priori power analysis estimated 225 as the minimum sample required to achieve 85% power to detect a small effect (Cohen’s w = 0.15) at an alpha of 0.05. Here and across all our studies, we oversampled to ensure more than adequate power. Participants were 32.66 years old in average (SD = 12.01), and 55 percent of them were female. The majority of participants had some college (28%) or college or higher education (56%), whereas the rest had some high school or a GED certificate (15%), or less than high school education (1%).In January 2021, participants were recruited from the Prolific platform (prolific.com), with quotas for US Non-Hispanic White  and entailed a 3-cell, within-subjects experimental design. A priori power analysis estimated 210 as the minimum sample required to achieve 85% power to detect a small effect (Cohen’s d = 0.20) at an alpha of 0.05. This design contained 3 conditions  manipulated within subjects, with each participant rotating through each condition presented in random order. Because one participant did not complete all intention measures, the final sample was N = 359. Participants were 50 percent female and 37.99 (SD = 11.10) years old on average. With respect to race, 79 percent of the respondents were white, 10 percent were black, 11 percent were Asian-American, 1 percent were American Indian or Alaska Natives, and 0.2 percent were Native Hawaiian or Pacific Islander. With respect to ethnicity, 12 percent were Hispanic.In March 2021, we recruited 360 participants recruited from Mechanical Turk (they require that you get vaccinated against a new disease. (Required Condition)You are about to begin a new job! The company tells you that they prefer that you get vaccinated against a new disease but that vaccination is up to you (Freedom of Choice Condition)You are about to begin a new job! The company tells you that give you more freedom and flexibility. (Control for Choice Freedom Condition).You are about to begin a new job! The company tells you that getting vaccinated against a new disease will Participants were told that they had a new job and the company told them (a) “they require that you get vaccinated against a new disease” (required condition), (b) “they prefer that you get vaccinated against a new disease but that vaccination is up to you” (preferred condition), and (c) “getting vaccinated against a new disease would give you more freedom and flexibility” (control for choice freedom condition). The conditions consisted of the following information (bold is ours):extremely unlikely) to 5 (extremely likely). Specifically, intentions were measured on a 1 to 5 scale, where 1 = extremely unlikely, 2 = somewhat unlikely, 3 = neither likely nor unlikely, 4 = somewhat likely, and 5 = extremely likely. This item was analyzed with parametric methods retaining all five points of the scale.The dependent measure was the intention to vaccinate, which was measured by the question, “How likely are you to get the vaccine?” with responses measured on a scale from 1 (https://www.mturk.com/) and the study involved a 2 (Norms: 70% vs. 30% support vaccine) × 3  within-subjects experimental design. This design contained 6 conditions  manipulated within subjects, with each participant rotating through each condition presented in random order. A priori power analysis estimated 276 as the minimum sample required to achieve 85% power to detect a small effect (Cohen’s d = 0.20) at an alpha of 0.05. Because one participant did not complete all intention measures, the sample used in the analyses of intentions was N = 357.In March 2021, we recruited 358 participants from Mechanical Turk (SD = 10.93) years old on average. With respect to race, 77 percent of the respondents were white, 12 percent were black, 12 percent were Asian-American, 0.3 percent were American Indian or Alaska Natives, and 0.3 percent were Native Hawaiian or Pacific Islander. With respect to ethnicity, 12 percent were Hispanic.Participants were 51 percent female and 37.48  that they require that you get vaccinated against the flu, COVID-19, and tetanus (requirement condition); (b) they prefer that you get vaccinated against the flu, COVID-19, and tetanus but that vaccination is up to you (freedom of choice condition); and (c) that being vaccinated against the flu, COVID-19, and tetanus will give you more freedom and flexibility within the job (control for choice freedom condition). These statements were presented to all participants twice accompanied with the statement that (a) You find out that 70% of their employees are vaccinated (higher norm condition) or (b) You find out that 30% of their employees are vaccinated (lower norm condition). Thus, as explained previously, six statements were randomized within subjects. That is, each participant read each statement, and after each one, participants were asked about their intentions to vaccinate with the question, “How likely are you to get these vaccines?” Responses were provided on a 1 to 5 scale, where 1 = 21. The items of the scale are (a) “It disappoints me to see others submitting to standards and rules,” (b) “I become frustrated when I am unable to make free and independent decisions,” (c)“When someone forces me to do something, I feel like doing the opposite,” (d) “I am content only when I am acting of my own free will,” and (e) “It makes me angry when another person is held up as a role model for me to follow.” Responses were provided on a scale from 1 (does not describe my feelings) to 5 (clearly describes my feelings). The internal consistency of the scale was adequate . This scale was analyzed with parametric methods retaining all five points of the scale.We also measured psychological reactance using a short scale adopted fromIn April 2021, participants were recruited from Qualtrics panels with nationally representative quotas for gender, education, and ethnicity. A comparison of demographics in our sample and the general US population appears in Table d = 0.20) at an alpha of 0.05.The design included two cells, one corresponding to the vaccination requirement and the other to freedom of choice. Participants were randomized to each condition of the between-subjects design. In addition to the quotas, Qualtrics ensures that participants proceed at a normal speed, thus removing people who complete the study by allocating less than 9 min in total. Qualtrics also selects out participants who do not pass their attention checks. A total of 606 participants were recruited with these procedures. A priori power analysis estimated 402 as the minimum sample required to achieve 85% power to detect a small effect (Cohen’s SD = 19.23), and 51 percent of them were female. The majority of participants had at least a high school education (87%), although 13% had less than high school education. Eighteen percent of participants had completed college and 14 percent had at least some postgraduate education. Thus, the sample clearly approximated the US population.Participants were 50.63 years old in average . Given that 126 million US inhabitants were employed full time in April 2021 out of 329.48, the rate of full time employment is similar. Considering that 12.1 percent might have been unemployed, this was similer to the US peak rate of 14.9 percent during the COVID-19 pandemicextremely unlikely, 2 = somewhat unlikely, 3 = neither likely nor unlikely, 4 = somewhat likely, and 5 = extremely likely. This item was analyzed with parametric methods retaining all five points of the scale.In this study, participants were randomly assigned to one or two scenarios: (a) On your job, the company tells you that they require that you get vaccinated against a new disease (requirement condition), or (b) On your job, the company tells you that you have to make the decision of vaccinating against a new disease for yourself (freedom of choice condition). They were then asked: “How likely are you to get the vaccine?” Their responses were provided on a scale on a 1 to 5 scale, where 1 = extremely unlikely) to 5 (extremely likely). These statements were used to create four indexes. An index of perceived benefits was created by averaging the following items: (a) “The vaccine will protect me from this new disease,” (b) “The vaccine will protect other people from this new disease,” (c) “The vaccine will allow me to go to work,” (d) “The vaccine will please my coworkers,” (e) “The vaccine will please my employer,” (f) “Getting the vaccine will allow me to fit in better,” and (g) “The vaccine may be risky” (reversed-scored) .  An index of perceived norms was created by averaging the following items: (a) “The majority of potential employees must want this vaccine,” (b) “Most employees are getting this vaccine,” and (c) “The majority of potential employees are probably getting this vaccine” . . An index of perceived fairness was built by averaging (a) “Getting the vaccine is fair” and (b) “Getting the vaccine respects me as a person” (inter-item r = 0.89). An index of perceived obligation was built by averaging (a) “Getting the vaccine is the moral thing to do” and (b) “Getting the vaccine is being a law-abiding citizen” (inter-item r = 0.86). These scales were analyzed with parametric methods retaining all five points of the scale.In addition, this study measured several potential mediators. Participants were told “When your employer tells you that you have to make the decision of vaccinating against a new disease for yourself, what do you think?” They then rated 18 statements on a scale from 1 (Supplementary Information.


Celiac disease (CD) is an immune-mediated enteropathy. CD may also involve complications with the oral cavity, which can result in various dental and oral pathologies. There are currently a limited number of studies on the oral manifestation of CD. This study aims to compare the oral manifestations of children with CD against healthy controls in Saudi Arabia.This study includes 208 children aged 6–14 years, distributed equally into CD patients and healthy controls. A parent completed and validated the interview questionnaire, which included the child's personal information and medical history. A dental examination was undertaken to measure possible recurrent aphthous stomatitis (RAS), dental enamel defects (DEDs), dental caries experience, and dental malocclusion. Data were analyzed using descriptive statistics and bivariate and multivariate analysis.P < 0.001)  and more DEDs than healthy controls  . No significant difference was found in the frequency of malocclusion between cases and controls.Two hundred and eight participants were included (104 CD patients and 104 controls). The mean age for CD patients was 10.67 ± 2.39 years and 10.69 ± 2.36 for the healthy controls. CD children had more RAS than controls (42.3% vs. 15.4%, Saudi Arabian children with CD had a greater number of clinical findings of RAS and DEDs than healthy controls. Pediatric dentists should consider the possibility of CD in child patients presenting with RAS or DEDs. Celiac disease (CD) is an immune related enteropathy in which small intestinal mucosal damage occurs in response to dietary gluten and other environmental triggers in individuals with genetic susceptibility; in particular, individuals carrying the HLA class II DQ2 and DQ8 haplotypes are at greatest risk .The possible clinical presentations of CD can be categorized into intestinal and extraintestinal manifestations. The classical intestinal manifestations include diarrhea, abdominal pain, abdominal distension, poor weight gain, and malnutrition . The nonRAS is usually described as an oval or rounded painful lesions, with well-defined margins that are surrounded by a reddish base. It can affect all pediatric age groups, and mostly involves the labial mucosa or the lateral surface of the tongue . RAS hasIn general, understanding of CD among many health care professionals, especially dentists, is minimal, so delayed diagnosis of CD is not unusual . To the This is a case–control study based on the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE)  guidelinThe sample size was calculated based on a previous study  using thThe inclusion criteria were the following for the patient group: children aged 6–14 years with biopsy-proven clinical diagnosis of CD based on the European Society for Pediatric Gastroenterology, Hepatology, and Nutrition (ESPGHAN) criteria [Patients were recruited in the pediatric gastroenterology clinic at King Abdulaziz University Hospital (KAUH). One hundred forty-nine contact numbers from the list of CD patients provided by the GI department were contacted; 104 subjects responded and agreed to participate (response rate was 69.8%). In order to select a control group, each CD subject was given two invitation letters of enrolment to distribute randomly to two classmates. A total of 208 invitation letters were provided for the CD children (the response rate was 50%). If parents approved their child’s participation, a phone contact was carried out to make sure that the participants were healthy. Parents were asked to bring their child to the Pediatric Dentistry Clinics in King Abdulaziz University Dental Hospital (KAUDH). The child and parents were taken to the dental clinic, where parents were asked to complete a validated parent questionnaire. This step was followed by a clinical examination of the child. The dental exam concerned RAS and DEDs using Aine's classification system . Caries The parent questionnaire was validated following its development. The validity was tested using content validity. It was given to a group of five medical and dental experts for the assessment of each item, based on specific criteria. Each item was given a score, and then a total average score was determined by dividing each count by the total number of experts. The content validity index for items (I-CVI) was calculated; any item with a score < 0.78 was modified in terms of structure or language. Following this, the content validity index for the whole scale (S-CVI) was calculated . The meaIn order to test the reliability of the parent questionnaire, 20 parents of children attending KAUH not included in the study were interviewed (in the Arabic language). The interview was repeated two weeks later for test–retest reliability; the test illustrated high reliability, with an intraclass correlation (ICC) of 0.82.t-tests, or analysis of variance (ANOVA) tests. Post-hoc analysis using Tukey correction was used in cases of statistically significant results in ANOVA tests. Following this, a multivariate analysis was carried out (as multiple logistic regression models) to predict the study outcomes, which were used to control for possible confounding between CD and other covariates. The choice of covariates in the models was based on the significance level in the bivariate analysis, i.e., all covariates that showed a significant association with the outcome were included in the model.All data were revised for completeness and logical consistency, then inserted into the Statistical Package for the Social Sciences  program, version 21, for analysis. Data entry was checked twice, then checking for duplicates was performed by SPSS, followed by examining each numerical variable for outliers using dot plot, and outliers were verified. Descriptive statistics were used to examine sample characteristics; continuous variables were summarized as means and standard deviations; and counts and percentages were used to categorical variables. To explore the association between CD and the primary outcomes , bivariate analyses were conducted with chi-squared tests, Fisher's exact tests, independent Ethical considerations were followed according to the Declaration of Helsinki. Approval to execute the study was obtained from the Research Ethics Committee at King Abdulaziz University (KAU) (number 078–09-17). Also, parents had to sign a parental consent form for the participation of their child.This case–control study comprised 208 children, aged 6–14 years, divided equally into 104 patients in the case group and 104 healthy children in the control group, all of which satisfied the inclusion and exclusion criteria.P = 0.971). In both the CD and control groups, 50% of participants were girls. Comparing the CD patients and control groups, there were no statistically significant differences between the two groups in terms of gender (P = 1.00), age categories (P = 0.980), family income (P = 0.874), mother education (P = 0.052), and father education (P = 0.781).Socio-demographic data revealed a mean age of 10.67 ± 2.39 years for the children with CD and 10.69 ± 2.36 years for the healthy controls (P = 0.037). There were no statistically significant differences between CD patients and the controls in terms of frequency of tooth brushing (P = 0.535), getting help with brushing (P = 0.889), swallowing toothpaste (P = 0.42), water source (P = 0.249), use of Zamzam holy water (P = 0.507), consumption of sweets (P = 0.323), intake of carbohydrates (P = 0.308), and daily intake of soft drinks (P = 0.369).The oral hygiene and dietary habits of the participants are shown in Table P < 0.001) , and CD children were 4.45 times more likely to have DEDs than healthy controls . Figure In this study, RAS was found in 42.3% of CD children and 15.4% of the controls (P = 0.019). Also, the mean number of decayed teeth in children with CD was significantly lower (3.70 ± 3.65) than healthy controls (5.13 ± 3.86) (P = 0.018). When considering the dental caries in the permanent dentition, DMFT showed no significant difference between the two groups (P = 0.200), but the mean number of decayed permanent teeth in children with CD (2.81 ± 3.21) was significantly higher than healthy controls (1.90 ± 2.18) (P = 0.021). Furthermore, when the dental caries consequences were analyzed as a continuous variable, the mean pufa in primary teeth was statistically lower in the CD group (1.44 ± 2.14) compared to the control group (2.66 ± 3.10) (P = 0.004). When the dental caries consequences were analyzed as a continuous variable, the pulpal consequences were significantly lower in children with CD (1.59 ± 2.05) than healthy controls (2.84 ± 3.19) (P = 0.005). Also, the pulpal consequences were significantly lower in children with CD (1.33 ± 1.93) than healthy controls (2.67 ± 3.11) (P = 0.001). The mean PUFA in permanent teeth was not statistically different between the two groups (P = 0.096) but the pulpal consequences were significantly higher in children with CD (0.46 ± 1.02) than healthy controls (0.22 ± 0.53) (P = 0.044).Dental caries experience among the study participants are shown in Table P < 0.001). Also, the CD group were 4.39 times more likely to have DEDs  than controls. CD was not shown to be a significant predictor of dental caries (P = 0.235) in this model  compared to children without CD (Celiac disease (CD) is an autoimmune enteropathy affecting the small intestine in which the ingestion of gluten, possibly with other environmental factors, trigger damage to the intestinal villi, often leading to atrophy . Oral maDEDs occur during the period of teeth formation when exposure to gluten may play a role in development . Gluten Several reports of the association of dental caries in children with CD have shown inconsistent results , 10–14. Perhaps the increased severity of decayed permanent teeth in the current study is related to DEDs that affect tooth structure, the lower calcium to phosphorus ratio, or the decreased salivary flow, resulting in more straightforward dissolution in CD patients .Our study found that dental caries experience as well as decayed primary teeth were lower in children with CD, which may be due to good adherence to a GFD, although adherence to a GFD was not assessed in the present study. Another study found that CD children, who were compliant with a GFD, had a lower number of caries than both those that were less compliant and controls , which mIn this study, to assess the caries experience, the DMFT/dmft and the PUFA/pufa indices were used, assessing diseases occurring pulpo-periapically, which could result from untreated dental caries. This index may help to understand the complete picture of caries present in an individual .We did not find any difference in the occurrence of malocclusion between children with CD and healthy controls. This finding contradicts one previous study that reported a higher rate of malocclusion in CD patients than in healthy controls . HoweverA strength of the present study is its case–control design, which involves the presence of an adequate sample size, homogeneity of the population studied, and the combination of clinical assessment and information gathered from a validated questionnaire.In conclusion, children with CD have higher odds of RAS and DEDs than healthy controls. Also, CD patients had a higher prevalence of dental caries in permanent teeth and a lower frequency in primary teeth.


We performed a dose–response meta‐analysis to estimate the association between lipid profiles with the risk of dementia and the potential differences according to the subtype of dementia based on prospective studies.We searched PubMed, Embase and Web of Science for relevant articles and performed a meta‐analysis. We applied fixed or random‐effects models to calculate pooled relative risk (RR) with their 95% confidence intervals (CI). The dose–response relationship was assessed by restricted cubic spline.Twenty‐five prospective studies comprising 362,443 participants and 20,121 cases were included in the final analysis. We found that increased risk of all‐cause dementia could be predicted by elevated total cholesterol (TC) . When looking at dementia subtypes, we also observed high TC and triglycerides (TG) may increase the future risk of Alzheimer's disease (AD), with a pooled RR of 1.13 (95% CI: 1.06–1.21) and 1.10 (95% CI: 1.04–1.15) respectively. Moreover, a dose–response analysis revealed a linear association between TC or TG and the risk of AD, with a pooled RR of 1.09 (95% CI: 1.02–1.16) and 1.12 (95% CI: 1.05–1.21) for per 3‐mmol/L increment in TC and TG, respectively.Current evidence suggest that every 3‐mmol/L increase in blood TC or TG is linearly associated with a 9% or 12% increase in RR of AD, supporting the notion that high TC and TG levels appear to play a causal role in the development of AD. Dementia comprises a heavy health and economic burden throughout the world, affecting 5–7% among those over 60 years and 80% among those over 90 years. Given the fact that cognitive impairment might take decades to develop and there is no cure available for dementia diseases, the wisest option would be to aim aggressively at prevention. The identification of a modifiable risk factor would represent a major advance in the prevention of dementia, but so far no single risk factor has been clearly discovered.Therefore, we performed an updated, more comprehensive meta‐analysis to estimate the association between individual components of lipid with the risk of incident dementia and the potential differences according to the subtype of dementia: AD or vascular dementia (VaD), as well as cognitive impairment without dementia (CIND) based on prospective studies. Furthermore, dose–response meta‐analyses were also conducted to explore the dose‐risk relationship with available data for them.This systematic review and meta‐analysis was prepared following recommendations of the Meta‐analysis Of Observational Studies in Epidemiology (MOOSE) and the Preferred Reporting Items for Systematic Reviews and Meta‐Analyses (PRISMA) groups.We carried out a literature search to March 2021 using the Pubmed, Embase, and Web of Knowledge databases without language restriction, using the search terms “lipids”, “cholesterol”, “triglycerides”, “low density lipoprotein”, “high density lipoprotein”, “dyslipidemia”, “hyperlipidemia”, “hypercholesterolemia”, “hypertriglyceridemia”; “dementia”, “Alzheimer”, “cognitive” and “MCI”; “prospective”, “cohort”, “follow up”, “longitudinal”. The reference lists of eligible articles and relevant reviews were hand‐searched for additional citations.Eligible studies had to meet the following criteria simultaneously: (1) were cohort study published in a peer‐reviewed journal; (2) participants were selected from general populations or memory clinic without dementia; (3) dementia or cognitive decline were recorded by use of well‐defined criteria; (4) reported adjusted risk estimates and 95% confidence intervals (CIs). For the dose–response analysis, the level‐specific case numbers and sufficient data for deriving these numbers were required. Blood lipid parameters  concentration were classified into three or more categories.The following data were extracted from each study by two authors (Zhu Y and Zhu RX): first author and publication year; region (or country); blood lipid parameters; gender distribution ; mean age or age range; follow‐up time; number of participants, number of cases ; outcome definition (relative risk [RR]/odds ratio [OR]/hazard ratio [HR]); covariates adjusted; levels of exposure, along with corresponding risk estimates and 95% CI for each category. For the study had several adjust models, we only included the result adjusted by the most covariates.The study quality was evaluated with the Newcastle‐Ottawa Quality Assessment Scale (NOS).In this meta‐analysis, all associations were estimated as RRs and 95% CIs; HRs or ORs is considered to mathematically approximate RR.A two‐stage, random‐effects, dose–response meta‐analysis was carried out taking into account the between‐study heterogeneity proposed by Orsini et al.I2 statistics, and a value of I2 above 50% or p < 0.05 indicated substantial heterogeneity.p < 0.05. In addition, we visually examined funnel plots for asymmetry. If possible, publication bias was indicated, we also used the trim‐and‐fill method to recalculate the pooled risk estimate.p‐values) were two‐sided, with p < 0.05 considered statistically significant.Heterogeneity was assessed by Cochrane's Q test and In brief, the search strategy returned 1102 records, from which we excluded 1041 after screening the title or abstract, leaving 59 potentially eligible publications. In addition, two eligible articles were identified through screening of reference lists. After reviewing the full‐text, 36 articles were further excluded. Finally, a total of 25 articles were included in the systematic review and meta‐analysis. Detailed information on the literature search could be seen from Figure All these studies published from 1995 through to 2020 with 362,443 participants and 20,121 cases of cognitive disorders were all prospective studies and the follow‐up period ranged from 1.5 to 40 years . We found no publication bias with Begg's and Egger's tests . Sensitivity analysis revealed that the results were identical to the primary results after removing any one of the studies one at a time. This indicated that no single study exerted a substantial influence on the pooled RR  and late‐life exposure (60+ years). In subgroup analyses according to age exposure, similar significant relationship between blood TC and all‐cause dementia risk was observed for studies confined to a midlife exposure  but not late‐life exposure  without significant heterogeneity (p = 0.39), and there was no linear dose‐risk relationship between them (p = 0.104) without substantial heterogeneity . We found no publication bias with Begg's and Egger's tests . In addition, the results of the sensitivity analysis were not substantially different. This indicates that our results are reliable . We summarized the detailed results in Table When we stratified studies by age exposure, similar significant relationship between blood TC and AD risk was observed for studies confined to a midlife  and late‐life exposure  without significant heterogeneity (p = 0.707). We found a linear dose–response association between TC level and risk of AD. For each 1‐mmol/L increase in TC level, the pooled RR of AD was 1.03 with little heterogeneity . The linear relationship was consistent among subgroup analysis by age exposure (midlife and late‐life), whereas the magnitude of association was stronger among midlife.Four studies were considered ineligible for inclusion in the dose–response analysis due to a lack of information regarding value of cases or participants, or provided TC levels for less than three categories.I2 = 0%; pheterogeneity = 0.491; Fig. p = 1.0, 0.894, respectively).Five prospective cohort studies including a total of 149,041 participants and 859 cases were included in the analysis comparing extreme categories.I2 = 71.1%; pheterogeneity = 0.016; Fig. p = 0.089, 0.088, respectively), the statistical power was too low to draw definitive conclusions regarding the association.Four prospective cohort studies including a total of 11,061 participants and 644 cases were included in the binary analysis.In addition, two reported data findings on TC and CIND that were incompatible with other studies for pooling. An early USA studyI2 = 38.8%; pheterogeneity = 0.201), the statistical power was too low to draw definitive conclusions regarding the association.Overall, four prospective cohort studies including a total of 17,996 participants and 1165 cases were used to explore the association between TG level and risk of all‐cause dementia Table .23, 3I2 = 27.8%; pheterogeneity = 0.245). We found no publication bias with Begg's and Egger's tests . Sensitivity analysis revealed that the results were identical to the primary results after removing any one of the studies one at a time. This indicated that no single study exerted a substantial influence on the pooled RR . We found a linear dose–response association between TG level and risk of AD. For each 1‐mmol/L increase in TG level, the pooled RR of AD was 1.04 with little heterogeneity .Three cohort studies were chosen for the dose–response analysis between TG and the risk of AD,I2 = 36.7%; pheterogeneity = 0.192; Fig. p = 0.308), but Egger's test indicated publication bias (p = 0.007), suggesting that the result should be interpreted with cautions.Four prospective cohort studies including a total of 15,939 participants and 266 cases were included in the analysis comparing extreme categories.Only two prospective cohort studies were found in the analysis to evaluate the relationship between LDL‐C levels and risk of all‐cause dementia.I2 = 40.7%; pheterogeneity = 0.134; Fig. p = 0.707, 0.639, respectively).Six prospective cohort studies including 191,554 participants and 11,861 cases were included in the binary analysis.I2 = 33.4%; pheterogeneity = 0.212). We found no publication bias with Begg's and Egger's tests .Four prospective cohort studies including a total of 9944 participants and 280 cases were included in the analysis comparing extreme categories.Only one prospective cohort study including a total of 864 participants and 334 cases was found in the analysis comparing categories of highest and lowest concentration of LDL‐C in late‐life.I2 = 59.7%; pheterogeneity = 0.03). We found no publication bias with Begg's and Egger's tests .Overall, six prospective cohort studies including a total of 19,363 participants and 1432 cases were used to explore the association between HDL‐C level and risk of all‐cause dementia Table .15, 2I2 = 49.7%; pheterogeneity = 0.037). We found no publication bias with Begg's and Egger's tests .Ten prospective cohort studies including 206,832 participants and 12,960 cases were included in the binary analysis.I2 = 68.4%; pheterogeneity = 0.023). We found no publication bias with Begg's and Egger's tests . But the statistical power was too low to draw definitive conclusions regarding the association.Four prospective cohort studies including a total of 17,598 participants and 351 cases were included in the analysis comparing extreme categories.I2 = 72.9%; pheterogeneity = 0.025). We found no publication bias with Begg's and Egger's tests .Three prospective cohort studies including a total of 2960 participants and 664 cases were found in the analysis comparing extreme categories.I2 < 50%), and high quality (NOS >6). Sensitivity analyses suggested that the primary pooled effect size of our analysis remained stable after the exclusion of any single study.We performed this systematic meta‐analysis based on 25 prospective studies involving 362,443 participants and 20,121 cases of cognitive disorders revealed a clear correlation of lipid parameters with the incident of dementia. The present study included only prospective studies with relatively large number of subjects. Therefore, our results represent the most recent information available on the associations between lipid parameters and risk of cognitive disorders. There are several key findings. We found the increased risk of all‐cause dementia could be predicted by an elevated blood concentration of TC. When looking at dementia subtypes, we observed distinct patterns for the different lipid fractions: higher baseline TC and TG concentrations may increase the future risk of AD, which is consistent with the previous data on the association between lipid levels and dementia risk.In the present study, we found that high TC levels increase the risk of all‐cause dementia and AD by 13%. Nevertheless, we observed different results in the association between TC and all‐cause dementia risk according to age. TC was an independent risk factors for all‐cause dementia in the midlife subgroup with 16% increased risk, whereas not to late‐life, in agreement with previous findings.A major finding of our study concerns TG and found that a significant association between TG levels and the risk of incident AD, suggesting increases of 10%. Moreover, our dose–response analysis revealed that for each 1‐mmol/L increase in TG levels, the risk of AD increased by 4%, for each 3‐mmol/L increase in TG levels, the risk of AD increased by 12%, respectively. Most previous studies found no significant associations of incident dementia  with TG.Results of a previous meta‐analysis evaluating the association between cholesterol and dementia were consistent with our study.The mechanism underlying the linkage between lipid and dementia risk remains speculative but is biologically plausible. High TC and TG indicate dyslipidemia (unfavorable lipid profiles) and were both detrimental to cognition in our meta‐analysis. It had been demonstrated that such a distribution of lipid levels was linked to multiple adverse outcomes, including arteriosclerosis or cerebrovascular diseases,Our meta‐analysis has several key strengths. First, we explored all comprehensive, high quality, available data from prospective studies reporting on the association of lipid profiles  with cognition on clinical status. Few previous studies have taken into account both lipid components and dementia type‐specific associations, including all‐cause dementia, AD, VaD, as well as CIND. We performed subgroup, sensitivity analyses which provided reliable and precise estimates. Second, as only prospective studies were included, we were able to predict the risk of dementia and cognitive decline before the disease onset. In addition, with respect to the strict inclusion criterion of providing information, our findings are not likely due to recall bias or selection bias. Consequently, and unlike in case–control studies, it is less likely to be confounded by indication. Third, the participants included in our study were mainly community‐dwelling elders selected from multiple sampling, so our findings could provide recommendations for this specific population. Moreover, most of the studies included in our analysis had a large sample size, a complete quality assessment, long follow‐up duration, and sensitivity and influence analyses, which provided high statistical power to our assessment of the correlation between lipid profiles and the risk of dementia and cognitive decline. Finally, in addition to comparing extreme categories of lipid, we also generated linear or non‐linear dose–response curves for TC and TG, which can help to quantify the associations and test the shape of these possible associations compared with previous meta‐analyses on this topic.Nevertheless, several limitations may affect the interpretation of our results. First, due to inherent limitations of the observational study design, we could not infer a causal relationship between dyslipidemia and dementia, including AD. Second, the adjusted covariates of each study are various and there should be residual confounding that cannot be adjusted. As for this, we include the RRs with the most adjusted covariates in this analysis. Third, meta‐analysis is a statistical method for quantitative synthesis of different study data. Thus, heterogeneity is inevitable in meta‐analysis, and it can also affect the interpretation of the results. The results with high heterogeneity should be interpreted carefully. Whereas, in our meta‐analysis we failed to find the source of heterogeneity. Fourth, multiple factors are associated with the concentration of lipid, and lipids themselves follow complex regulatory pathways. Although maximally adjusted estimates were applied, over‐ or under‐adjustment may exist and cofound the results. Finally, we could not conduct the dose–response analysis to test all of the hypothesis because of the actual number of studies was insufficient to draw a firm conclusion among partial subgroups.Taken together, the alteration in blood concentration of lipid profiles precedes the incidence of dementia, though the related etiology is a subject of ongoing research and debate. Throughout life, genetic and environmental  factors have interactive effects in predisposing a person to dementia. If high TC or TG level is an important risk factor for AD, it offers the possibility for the prevention of the disease via diet, exercise, and medication as recommended or required. However, it should be noted that the effect of controlling dyslipidemia may be stronger in early AD stages or mild cognitive disorders. Once the pathological change has progressed enough to require medication, prevention efforts could not reach significant effects.In conclusion, the results of our meta‐analysis indicate that lipid profiles of high TC and TG were significantly correlated with an elevated the risk of AD. Particularly importantly, there was a linear, dose‐dependent relationship between TC or TG levels and risk of incident AD. Therefore, we recommend greater attention be paid to cognitive besides metabolic effects when the concentration of lipid profiles in blood deviates from normal. Modification of high TC and TG even late in life may reduce not only the risk of cardiovascular disease but also the most prevalent age‐related neurodegenerative disease, AD. In the future, more well‐designed RCTs with long‐term follow‐up and serial assessments of lipid are needed to further clarify the causal relationship between lipid and dementia.The authors declare no conflict of interest.Table S1. Characteristics of the prospective cohort studies included in the meta‐analysis.Figure S1. Dose–response relationship between TC and risk of all‐cause dementia.Figure S2. Overall pooled analysis of association between TC levels and VaD.Figure S3. Overall pooled analysis of association between TC levels and CIND.Figure S4. Overall pooled analysis of association between TG levels and VaD.Figure S5. Overall pooled analysis of association between LDL levels and AD.Click here for additional data file.


Several studies have investigated the effects of different medications on orthodontic tooth movement (OTM). This study assessed the effect of caffeine injection on OTM in rats.  Thirty-five male Wistar rats were randomly divided into five groups. Their first molars and central incisors were attached with a nickel-titanium closed coil spring with 50 g load. The rats in the three experimental groups received 25, 50, and 75 mg/kg caffeine intraperitoneally for 21 days. The negative control group did not receive any injection and did not undergo orthodontic treatment. The positive control group underwent orthodontic treatment and received 0.9% NaCl  injection. After 21 days, all rats were sacrificed by chloroform inhalation, and the maxilla was resected. The mean number of Howship's lacunae, blood vessels, osteoclasts, and root resorption lacunae was histologically measured. The bone volume-to-total volume ratio (BV/TV) in maxillary molars was calculated by microcomputed tomography (micro-CT) to quantify bone loss.  The amount of OTM and the number of osteoclasts, blood vessels, and Howship's lacunae significantly increased in rats under caffeine therapy, while the number of root resorption lacunae did not increase. Lower BV/TV in the caffeine groups was in accord with the increased count of osteoclasts.  Caffeine injection can significantly increase OTM in rats. Orthodontic treatment has gained wide public acceptance in the recent years. It not only corrects dental anomalies but also improves the esthetic appearance of the face. Moreover, orthodontic treatment corrects temporomandibular joint disorders due to malocclusion . OrthodoCaffeine  is among the most common central nervous system stimulants used in drinks, foods, and medications , 15. CafThe reports regarding the effects of caffeine on bone metabolism, bone density, and bone regeneration have been controversial. Some authors reported that caffeine caused osteoporosis and periodontal disease , 18, 19,Evidence shows that the genome of rats and humans is highly similar such that approximately 90% of the human genes have direct orthologs in rats . Thus, wn = 7) [The study was approved by the ethics committee of Kermanshah University of Medical Sciences (IR.KUMS.REC.1397.724) and conducted in accordance with the ARRIVE guidelines . This ann = 7) , 27 as fGroup 1 received intraperitoneal injection of 25 mg/kg caffeine at the aforementioned time points for 3 weeks simultaneously with orthodontic treatment.The rats in group 2 received intraperitoneal injection of 50 mg/kg caffeine at the aforementioned time points for 3 weeks simultaneously with orthodontic treatment.The rats in group 3 received intraperitoneal injection of 75 mg/kg caffeine at the aforementioned time points for 3 weeks simultaneously with orthodontic treatment.The negative control group did not receive any injection and did not undergo orthodontic treatment. Caffeine was purchased from Sigma-Aldrich .The rats in the positive control group received intraperitoneal injection of 0.9% NaCl at the aforementioned time points for 3 weeks simultaneously with orthodontic treatment.2 for 10 s followed by the application of Transbond XT composite .The rats were anesthetized with 10% ketamine hydrochloride  and 2% xylazine . After anesthesia induction, the vital signs of the rats were closely monitored. Also, they were rotated from side to side every couple of minutes to prevent pulmonary edema. The room temperature was also controlled. The nickel-titanium closed coil springs were installed  to induce OTM. The first molars and central incisors were connected by a stainless steel wire. Next, the first molars and central incisors were etched with 37% phosphoric acid  for 30 s, rinsed for 10 s, and dried with an air spray for 15 s. Single Bond  was applied on the surface and light-cured with a LED curing unit  with a light intensity of 150 mw/cmThe load applied by the coil springs was 50 g  Figure . The magThe distance between the enamel of the most distal part of the first molar crown and mesial surface of the second molar was measured by an examiner blinded to the group allocation of rats at baseline and after 21 days. ABViewer 14 software was used for this purpose. Each measurement was repeated 3 times, and the mean of the measurements was reported as the final value. Impressions were made from the teeth at 1 and 21 days using injection silicone polyvinyl siloxane impression material . After 4 min , the impressions were checked and poured with dental stone . After 24 h, the gypsum cast was removed. The casts were scanned with a 3D scanner  to create 3D models in an STL format  Figure . Next, tµm thickness [After 21 days, the rats were sacrificed by inhalation of chloroform in a desiccator. The maxilla was resected and sent for histological analysis. The specimens were fixed in 10% formaldehyde and decalcified in 10% formic acid for 48 h . After fixation, decalcification of the tissue was performed using 12.5% ethylenediaminetetraacetic acid, and fixation was performed by using a fixator for 10 weeks. The decalcifying solution was agitated 10 times a day and refreshed twice weekly until complete decalcification occurred. Next, they were dehydrated by using ethanol and embedded in paraffin blocks. Parasagittal sections were made by using a microtome  with 5 hickness . An exam2. Each specimen was assessed three times, and the mean of the three measurements was recorded.Tissue specimens were stained with hematoxylin and eosin, and the slides were inspected by an experienced pathologist who was blinded to the group allocation of specimens under a light microscope  at 100x magnification. The number of Howship's lacunae, blood vessels, osteoclasts, and root resorption lacunae was also counted in an area measuring 0.01 mmµA, respectively, and frame exposure time was set to 1 second by 1.4 magnification. Total scan duration was 28 minutes. Slice thicknesses of reconstructed images were set to 50 micrometers. All the protocol settings were controlled by LOTUS-inVivo-ACQ software. The acquired 3D data were reconstructed using LOTUS-inVivo-REC by a standard Feldkamp, Davis, and Kress (FDK) algorithm. Also, LOTUS-inVivo-3D was used for rendering of reconstructed images, and by adding the bone analysis plugin (BAP) inside software, we reported bone volume (BV) and total volume (TV) parameters [In this study, we used an in vivo X-ray microcomputed tomography (micro-CT) scanner. LOTUS-inVivo has a cone beam microfocus X-ray source and a flat panel detector. In order to obtain best possible image quality, the X-ray tube voltage and its current were set to 60 kV and 130 rameters .t-test.The data were analyzed using SPSS version 21  by one-way ANOVA, Tukey's test, and The minimum intraclass correlation coefficient was calculated to be 0.964 for all variables, indicating excellent reliability of the measurements.P < 0.05). The amount of OTM was minimum in the negative control (P = 0.011) and maximum in the 75 mg/kg caffeine group (P= 0.458). A significant difference existed between the negative control group and all other groups in OTM (P < 0.05). Also, significant differences existed between the positive control and 50 mg/kg caffeine and positive control and 75 mg/kg caffeine groups (P < 0.05). However, the difference between the positive control and 25 mg/kg caffeine group was not significant (P = 0.354). Increasing the caffeine dosage significantly increased the mean amount of OTM such that the difference between the 25 and 50 mg/kg (P = 0.005) and 25 and 75 mg/kg (P < 0.05) caffeine groups was significant. However, the difference between the 50 and 75 mg/kg caffeine groups did not reach statistical significance . The number of Howship's lacunae was maximum in the 75 mg/kg caffeine group (P = 4.166) and minimum in the negative control group (P = 0.166). Significant differences existed between the negative control and all other groups in the number of Howship's lacunae (P < 0.05). However, the difference was not significant between the positive control and 25 mg/kg caffeine groups (P = 0.540). The difference between the positive control and 50 mg/kg caffeine group was not significant either (P = 0.323). However, the difference was significant between the positive control and 75 mg/kg caffeine group (P < 0.05). The difference between the 25 and 50 mg/kg caffeine groups was not significant (P = 0.995), but the difference between the 25 and 75 mg/kg caffeine groups was significant (P = 0.014). Also, 50 and 75 mg/kg caffeine groups had a significant difference in this respect (P = 0.035).At 21 days, a significant difference existed in the number of Howship's lacunae among the five groups (P < 0.05). The number of blood vessels was maximum in 75 mg/kg caffeine (P = 5.6) and minimum in the negative control group (P = 2.0). Also, the difference between the negative control group and the other groups was significant (P < 0.05). No significant difference was noted between the positive control and 25 mg/kg caffeine (P = 1.000) or positive control and 50 mg/kg caffeine (P = 0.340) groups. However, the difference between the positive control and 75 mg/kg caffeine group was significant (P ≤ 0.001). The difference between the 25 and 50 mg/kg caffeine groups was not significant (P = 0.340), but 25 and 75 mg/kg caffeine groups (P ≤ 0.001) and 50 and 75 mg/kg caffeine groups had significant differences in this regard (P = 0.002).At 21 days, a significant difference was noted in the number of blood vessels among the five groups (P < 0.05). The number of osteoclasts was maximum in 75 mg/kg caffeine (P = 4.000) and minimum in the negative control group (P = 0.833). A significant difference existed between the negative control and the other groups in the number of osteoclasts (P < 0.05). No significant difference existed between the positive control and 25 mg/kg caffeine group (P = 0.139). However, the difference between the positive control and 50 mg/kg caffeine group was significant (P = 0.025). A significant difference was also noted between the positive control and 75 mg/kg caffeine group (P = 0.010). No significant difference was found between the 25 and 50 mg/kg (P = 0.925), 25 and 75 mg/kg (P = 0.742), or 50 and 75 mg/kg caffeine groups (P = 0.994).A significant difference existed among the five groups in the number of osteoclasts (P < 0.05). The number of root resorption lacunae was maximum in 50 and 75 mg/kg caffeine (P = 1.5) and minimum in the negative control group (P ≤ 0.01). No root resorption lacunae were seen in the negative control group, and this group had significant differences with all other groups in this respect (P < 0.05). No significant difference was noted between the positive control and 25 mg/kg caffeine (P = 0.672), positive control and 50 mg/kg caffeine (P = 0.870), positive control and 75 mg/kg caffeine (P = 0.672), 25 and 75 mg/kg caffeine (P = 0.870), or 50 and 75 mg/kg caffeine (P = 1.000) groups.At 21 days, a significant difference was noted in the number of root resorption lacunae among the five groups (P < 0.05). The BV/TV ratio was minimum in the 75 mg/kg caffeine and maximum in the negative control group. The negative control group had significant differences with the other groups in this regard (P < 0.05). No significant difference was found between the positive control and 25 mg/kg caffeine (P = 0.395) and positive control and 50 mg/kg caffeine (P = 0.230) groups. However, the difference between the positive control and 75 mg/kg caffeine group was significant (P = 0.031). Pairwise comparisons of the three caffeine groups showed no significant difference (P > 0.05, The micro-CT results revealed a significant difference in this regard among the five groups (Several studies have assessed the effects of different medications and surgical procedures on OTM. Some studies reported that certain medications or surgical approaches enhanced OTM, while some others showed deceleration of OTM by different stimuli –36. RespThis study assessed the effect of caffeine on OTM in rats. Rats are suitable laboratory animals to study bone remodeling and OTM in response to mechanical forces –42. ThisThe results of the present study revealed that orthodontic forces stimulated the osteoclasts and angiogenesis. Also, injection of 25, 50, and 75 mg/kg caffeine along with orthodontic force application further increased the number of osteoclasts and blood vessels. Although the difference in this respect was not significant among different doses of caffeine, caffeine injection eventually increased OTM. Increasing the caffeine dosage increased OTM. Histological assessments revealed that injection of caffeine increased the number of osteoclasts, Howship's lacunae, and blood vessels. However, it did not significantly increase the number of root resorption lacunae. The increase in the number of osteoclasts due to caffeine injection was in agreement with a previous report in this regard . Micro-CIn this study, all measurements were made by the same examiner to eliminate interobserver errors, which was a strength of this study.This study had some limitations. It was an animal study with limited sample size. The findings of animal studies should be further confirmed by clinical human studies. Ethically, a minimum acceptable number of laboratory animals should be used in animal studies, which limits the generalizability of the results. Human studies with larger sample size are required to assess the effect of caffeine on OTM over long courses of treatment.This animal study indicated that caffeine can accelerate OTM in rats, and this effect was fortified by increasing the caffeine dosage. Further investigations are required to elucidate the mechanisms involved in this effect. Generalization of the results of animal studies to humans should be done with caution. Given that the results are confirmed, the skeletal effects of caffeine should also be taken into account by orthodontists in orthodontic treatment planning.


The PAC determined by the CLEIA was significantly correlated with that measured by liquid chromatography/mass spectrometry or conventional radioimmunoassay. The ARC determined by the CLEIA was significantly correlated with that measured by immunoradiometric assay. The limits of detection of the CLEIAs for PAC and ARC were 0.1 ng/dl and 0.04 pg/ml, respectively, which were better than those of conventional methods . The PAC and PAC/ARC ratio (ARR) were significantly higher, and the ARC significantly lower, in patients with PA than in those with essential hypertension. An ARR cut-off of 1.31 ng/dl per pg/ml showed a sensitivity of 96.2% and specificity of 78.3% for PA screening. The newly developed CLEIAs for measuring PAC and ARC could provide a clinically powerful alternative to conventional methods used for hypertension screening in clinical practice.Determination of plasma aldosterone concentrations (PAC) and plasma active renin concentrations (ARC) is essential for the diagnosis of primary aldosteronism (PA). In Japan, although PAC and ARC are measured by radioimmunoassay and immunoradiometric assay, respectively, non-radioisotopic methods with better detection sensitivity, measurement accuracy, and technical simplicity are needed. We developed two-site sandwich chemiluminescent enzyme immunoassays (CLEIAs) to measure both PAC and ARC using monoclonal antibodies immobilized onto ferrite particles. The results of both assays are obtained simultaneously from a single plasma sample within 30 min using a fully automated system. The novel CLEIAs were validated using plasma samples from patients with PA ( Primary aldosteronism (PA) is a major cause of secondary hypertension –5. ConsiWe developed new CLEIAs for detecting PAC and plasma ARC characterized by high sensitivity, sufficient traceability back to the certified reference materials, and versatile and efficient implementation by an automated system.Patients referred to Kyoto Medical Center for further investigation of hypertension were evaluated. Diagnostic criteria for diagnosing PA is shown in Supplementary. In short, PA was diagnosed if the screening test was positive and at least one confirmatory test showed positive results. EH was diagnosed if any secondary cause of hypertension was excluded. The numbers of patients with PA, EH, and other miscellaneous diseases were 52, 23, and 33, respectively. Other miscellaneous diseases include renovascular hypertension, non-functioning adenoma, pheochromocytoma, paraganglioma, ACTH-independent macronodular adrenal hyperplasia, subclinical Cushing syndrome, and adrenocortical carcinoma were used for method comparison using peripheral blood. Apart from peripheral blood, we also evaluated the level of PAC and ARC by RIA and CLEIA using 50 samples from 5 patients for adrenal vein blood, and 18 samples from 18 patients for urine, respectively.n = 50) and urine samples collected from patients with hypertension (n = 18) were analyzed.As part of routine clinical investigations, blood samples were collected into EDTA-2Na tubes from the antecubital veins of patients in the sitting position for at least 30 min in the clinic. Plasma samples were stored at less than −20 °C and subjected to our CLEIA immediately after thawing to room temperature in a 20 °C water bath. In addition, plasma samples collected from the adrenal veins of patients with PA .® Presto II and LUMIPULSE® L2400) with CLEIAs enables simultaneous determination of PAC and ARC, respectively, from a single plasma sample, requiring ~20 min.Our CLEIAs for measuring PAC and ARC were developed using two monoclonal antibodies against different epitopes of aldosterone  and renin , respectively. The limit of detection (LoD), accuracy, precision, linearity, and recovery were determined for each CLEIA according to the recommendations of the Clinical and Laboratory Standards Institute . In addition, using the dedicated fully automated systems  against different epitopes of aldosterone. The plasma sample (30 µl) was incubated with 250 µl reagent A containing the primary antibody immobilized onto ferrite particles and incubated for 10 min at 37 °C. After washing with magnetic beads, the sample was incubated with 250 µl reagent B containing the ALP-conjugated secondary antibody and incubated for 10 min at 37 °C. After washing the magnetic beads, the complex consisting of the primary antibody, ALP-conjugated secondary antibody, and ferrite particles was dispersed, and 200 µl solution containing 3-(2′-spiroadamantane)-4-methoxy-4-(3″-phosphoryloxy) phenyl-1, 2-dioxetane disodium salt (AMPPD) was added to measure chemiluminescence . The uriFor ARC detection, we developed two-site sandwich immunoassay using two specific mouse monoclonal antibodies against different epitopes of renin. An anti-active renin mouse monoclonal antibody that binds to the active site of active renin specifically was used as the detection antibody. An anti-renin monoclonal antibody that recognizes the non-active site of renin was used as the capture antibody. The plasma sample (40 µl) was incubated with 250 µl reagent A containing the detection antibody immobilized to ferrite particles and incubated for 10 min at 37 °C. After washing with magnetic beads, the sample was incubated with 250 µl reagent B containing the ALP-conjugated capture antibody and incubated for 10 min at 37 °C. After washing the magnetic beads, the complex consisting of the detection antibody, ALP-conjugated capture antibody, and ferrite particles was dispersed, and 200 µl solution containing AMPPD was added to measure chemiluminescence. The CLEIA for ARC was calibrated using in-house human recombinant renin  as the reference material. The accuracy, precision, linearity, and recovery of the CLEIA were measured.Correlation and linear regression analyses were conducted to compare PAC measurements by our novel CLEIA with those by LC-MS/MS and a conventional RIA , and to compare plasma ARC measurements by our novel CLEIA with those by a conventional immunoradiometric assay (IRMA)  and PRA by a conventional RIA , respectively.PAC, ARC, and the PAC/ARC ratio (ARR) determined by the CLEIAs were compared between patients with PA and patients with EH.U test. In the method comparison study, regression analysis was performed using the Passing–Bablok method and Pearson’s correlation coefficients. Bland–Altman analysis was used to evaluate mean differences. The statistical significance was set at P ≤ 0.05. Statistical analyses were performed using Analyse-it software  and SAS for Windows .PAC, ARC, and ARR were compared between the patient groups using the Mann–Whitney Three different concentrations of certified reference materials for ensuring the traceability of aldosterone (NMIJ CRM 6402-a) were assayed in duplicate using our CLEIA for PAC. The certainty of the CLEIA for measuring the certified reference materials ranged from 99 to 104%   . To validate the CLEIA measurements, the PAC of two discrepant samples A and B in Fig. The PAC in the peripheral blood measured by the CLEIA was significantly correlated with the PAC measured by LC-MS/MS (41) Fig. . There w41) Fig. . Althoug32) Fig. , the cory = 0.84x − 51.2, r = 0.869, p < 0.01; n = 50)  Fig. . The PAC50) Fig. .Fig. 2Coy = 1.0x − 0.31, r = 1.00, p < 0.01; n = 50)   Fig. . The mea50) Fig. . The ARC53) Fig. .Fig. 3CoThe CLEIA PAC and plasma ARC measurements were compared between the patients with PA and those with EH. The baseline clinical characteristics of the patients are shown in Table The ARR measured by the CLEIA was significantly higher in patients with PA (3.31 ng/dl per pg/ml) than in patients with EH (0.52 ng/dl per pg/ml) Fig. . ReceiveIn this study, the conversion formulas between conventional and the new assays were as follows: PAC (CLEIA) = 0.62 × PAC (RIA) − 1.98, and PAC (CLEIA)/PRA (RIA) = − 1.05 + 0.49 × PAC (RIA)/PRA (RIA). For example, the ARR cut-off value of 20 determined by RIA for PAC and PRA, which is used to screen for PA, can be converted to 8.75 by PAC (CLEIA)/PRA (RIA). The PAC cut-off value of 6 determined by RIA, which is used to diagnose PA by the saline infusion test, can be converted to 1.74 by PAC (CLEIA).The prevalence of PA has reached 5–10% in the hypertensive population , 17, 18.These findings agree with previous reports of the advantages of CLEIAs –25. JapaOur CLEIA for measuring the plasma ARC demonstrated a good correlation with conventional assays. The CLEIA showed a detection sensitivity of 0.04 pg/ml, which was significantly better than that (5 pg/ml) of IRMA. The detection sensitivity of the ARC is important for determining the ARR, which is universally used for PA screening in patients with hypertension –10. The The present CLEIA is applicable as a standard assay method for measuring the PAC in daily clinical practice, and an automated system with these CLEIAs enables us to obtain PAC and plasma ARC results within 30 min. This significant decrease in measurement time will help improve the efficiency of the PA diagnostic process in the clinic and the success rate of adrenal venous sampling. The novel CLEIAs for PAC and ARC could affect the clinical practice of hypertension and PA.The present study has several limitations. First, blood sampling in the patients was not optimized in terms of food intake, posture, time of day, or use of antihypertensive medications, which may affect PAC and/or ARC measurements. However, the major aim of the present study was to validate the new CLEIAs and compare their efficacies with those of conventional assays. The sampling conditions likely did not affect the correlations between the assays. Second, an ARR cut-off in the current study is calculated by small sample size. Further systematic studies using larger numbers of samples obtained under standardized conditions are needed to establish the appropriate cut-offs of PAC and ARR for screening PA.Radioimmunoassay (RIA) has been used to determine the PAC, ARC, and PRA in clinical practice, although various issues with RIAs, such as the usage and disposal of radioisotopic materials, complexity of the manual assay, poor traceability of certified reference materials, and low detection sensitivity at lower concentrations, remain to be resolved.The novel CLEIAs using a fully automated system developed to measure PAC and ARC were characterized by a better detection sensitivity, much shorter measurement time, and lower cost compared with the conventional RIA. These new methods are expected to facilitate the diagnostic process and improve the quality of hypertension and PA clinical practice.Supple Figure 1SUPPLEMENTAL METHODS


Maclura tinctoria (Tajuva) heartwood. NFC-based aerogels were produced by freeze followed by lyophilization, in a way that the Tajuva extract acted as a functionalizing agent. The obtained aerogels showed high porosity (ranging from 97% to 99%) and low density (ranging from 0.025 to 0.040 g·cm−3), as well a typical network and sheet-like structure with 100 to 300 μm pores, which yielded compressive strengths ranging from 60 to 340 kPa. The reached antibacterial and antioxidant activities, percentage of inhibitions and water uptakes suggest that the aerogels can be used as fluid absorbers. Additionally, the immobilization of the Tajuva extract indicates the potential for dentistry applications.Aerogels are 3-D nanostructures of non-fluid colloidal interconnected porous networks consisting of loosely packed bonded particles that are expanded throughout their volume by gas and exhibit ultra-low density and high specific surface area. Cellulose-based aerogels can be obtained from hydrogels through a drying process, replacing the solvent (water) with air and keeping the pristine three-dimensional arrangement. In this work, hybrid cellulose-based aerogels were produced and their potential for use as dressings was assessed. Nanofibrilated cellulose (NFC) hydrogels were produced by a co-grinding process in a stone micronizer using a kraft cellulosic pulp and a phenolic extract from  One of the main technological fields in pharmaceutical research deals with the production of new multifunctional materials for the immobilization and controlled release of drugs according to the severity of each targeted injury. For Gustaite and co-workers , unlike Plant derivatives, such as (micro/nano)fibrillated cellulose (NFC), have been used in biomedical engineering due to their low toxicity, biodegradability, and biocompatibility ,4. NFC i2·g−1, a porosity of 84.0–99.9%, and good biodegradability  using commercial software OriginPro® 2017, as described by Missio and co-workers [−3. Morphology was evaluated using scanning electron microscopy  at a voltage of 10 kV. Cylindrical specimens with the dimensions of 33 × 20 mm (diameter × length) were tested for compression . For that, a pre-load of 0.05 N was followed by a compressive load applied at a 10 mm × min−1 crosshead speed until reaching 50% strain. Water uptake was evaluated by the percentage mass gain of prismatic specimens (with dimensions of 15 × 5 × 3 mm), which were immersed in distilled water until reaching constant mass. Prior to being weighted, excessive moisture of each sample was carefully removed using a filter paper. FTIR-ATR spectra  by means of triplicate readings of absorbance bands at 765, 510, and 517 nm, respectively. In the case of antioxidant capacity, the DPPH method was used, in which the aerogels were kept at room temperature for 24 h in total darkness. DPPH concentration was calculated using the following regression equation: Trolox concentration = 0.0045·corrected absorbance inhibition (in % − 0.0118). Gallic acid (0–80 mg·L−1), catechin (0–200 mg·L−1), and Trolox (0–0.250 mM) were used as standard solutions for preparing calibration curves to measure the contents of total phenolic, total flavonoids, and antioxidant capacity, respectively.For the following spectroscopic analyses, the bio-based aerogels were cryofracted using liquid nitrogen and extracted in an 80% ethanol aqueous solution under mechanical stirring  for 120 min and subsequent centrifugation . The supernatant phase was removed, and the final extract was diluted in a 10% ethanol solution. Phenolic compounds and flavonoids contents, as well as antioxidant capacity, were determined in vitro by UV-VIS spectroscopy  is promising for treatments in the oral cavity.-workers , who invThe antimicrobial activity of the TA extract is probably ascribed to its high content of phenolic compounds, such as tannins and flavonoids. In this context, the mechanism of action of phenolic compounds on pathogenic bacteria is not yet well known. According to Machado and co-workers , tanninsThe hybrid hydrogel incorporated with alginate presented a slightly smaller diameter if compared to that neat NFC-based one . AccordiThe IPD of both studied hydrogels were about 0.5, which indicates a good distribution of the NFC in the aqueous medium with neither large particles nor agglomerates . In fact−1 (ν(C–C) stretching vibration), 1050 cm−1 (ν(C–O) stretching vibration), 1240 and 1365 cm−1 (ν(C–H) bending), and 1730 cm−1 (ν(C=O) stretching of COOH). The TA extract showed several prominent bands in the 1000–1500 cm−1 range, which are attributed to C–O vibrations of (i) –C–O–C– ether bridges in complex oxygen-containing cyclic molecules and/or (ii) =C–O–C= breathing in furanic moieties, as well as in-plane C–H bending vibrations from aromatic molecules from heteroaromatic compounds [−1 is typically attributed to C=C stretching in the benzene ring. Finally, that broadband at 3250 cm−1 can be ascribed to stretching vibration of the O–H groups associated with absorbed moisture [−1 which are attributed to glycoside bond (C–O–C), symmetric stretching of –COO−, and antisymmetric stretching of –COO−, respectively [ompounds . That bamoisture . The FTIectively .−1 represent –C–H stretching, aromatic skeletal vibration, –C–O vibration, –OH bending, –C=O–linkage, –CH– stretching, and OH–stretching, respectively [Regarding the aerogels B, there ectively .−3 and porosities between 97% and 99%. The aerogel density may be attributed to the density of its cellulosic raw material [material  and, becmaterial . Therefomaterial .The entangled NFC fibres formed a free-standing 3D structure as shown in All the aerogels showed ductile curves A, which That stiffest aerogel (ACNE) presented the highest volume retention after the compression and the smallest water uptake . This beMacrosphyra Longistyla [Rhizophora racemose [Arbutus unedo [The contents of total phenolic compounds and flavonoids of the aerogels and their isolated phases are presented in ngistyla , Rhizophracemose , and Arbus unedo . Accordius unedo , those bThe TA extract also showed a high antioxidant activity, which was also conferred to the ACNE and ACNAE aerogels in a minor intensity. This antioxidant activity can be associated with the presence of phenolic and tannin compounds ,46. ThisThe Tajuva extract showed a high antioxidant activity probably ascribed to substances present in the extract, such as phenolic compounds, especially the tannins and other flavonoids. Enzyme inhibition assays provided indications on the potential therapeutic value of the Tajuva extract for the management of epidermal hyperpigmentation and neurodegenerative complications. However, further research including clinical in vivo studies is recommended to further evaluate these aforementioned properties, in order to include this traditional plant as a potential cosmetic, nutraceutical, and pharmacological ingredients. −3 and a porosity of 97–99% were marked by 100 to 300 μm pores distributed in a typical network and sheet-like structure. The aerogels presented good compressive strengths ranging from 60 to 340 kPa and good flexibility and durability and seem to be reusable, since they were capable of absorbing a large volume of water even after the compressive tests.Hydrogels incorporated with NFC, sodium alginate, and TA extract were successfully produced by the cogrinding process and showed good stability, as well as a uniform particle size distribution. The aerogels with a density of 0.025–0.040 g·cm