Abstract
Background
Measles elimination remains elusive in many countries, and resurgences occur even in previously controlled settings. We assessed age-specific clinical and epidemiological factors associated with laboratory-confirmed measles among suspected cases.
Methods
In this test-negative case–control study, we analyzed surveillance data from Diyarbakır, Türkiye (2012–2024). Cases and controls were defined by measles-specific IgM results. Cases were stratified into two age groups, individuals aged ≤ 1 year and individuals aged > 1 year, with controls selected using systematic sampling at a case-control ratio of 1:3. Univariable and multivariable logistic regression analyses estimated odds ratios (ORs) for demographic factors, vaccination status, seasonality, epidemic year, clinical symptoms, and hospitalization.
Results
Among 3,868 individuals (967 cases, 2,901 controls), 32% of cases were ≤ 1 year of age. In infants, predictors included hospitalization (OR: 7.95; 95% CI:4.21–15), upper respiratory tract symptoms (OR: 3.20; 95% CI:1.87–5.48), age 7–9 months (OR: 4.16; 95% CI:2.33–7.41) and 10–12 months (OR: 5.28; 95% CI:3.03–9.20), and spring–summer season (OR: 2.08; 95% CI:1.14–3.82). Among individuals aged > 1 year, significant predictors were receipt of a single dose (OR: 3.64; 95% CI:1.09–12.15); being unvaccinated (OR: 16.13; 95% CI:4.77–54.54) or having unknown vaccination status (OR: 17.26; 95% CI:5.14–57.88); epidemic year (OR: 4.06; 95% CI:2.74–6.02); spring–summer season (OR: 1.80; 95% CI:1.21–2.69); fever (OR: 1.89; 95% CI:1.20–2.95); and hospitalization (OR: 2.24; 95% CI:1.39–3.62).
Conclusion
Among suspected cases, age-specific predictors of laboratory-confirmed measles may support clinical suspicion, surveillance alertness, and prioritization of laboratory testing when interpreted alongside the local epidemiological context, including seasonality, vaccination coverage, and individual vaccination status. These findings may help guide regionally tailored surveillance and control efforts.
Keywords: Measles, Infants, Predictors, Public health, Vaccination
Background
Measles is one of the most contagious vaccine-preventable diseases, and achieving its elimination remains a global public health priority. Despite substantial reductions in the global measles burden since the early 2000s, measles cases were estimated to have increased by 8% in 2024 compared with pre-pandemic 2019 levels [1]. Consequently, measles continues to be a priority area requiring strengthened approaches to clinical diagnosis, surveillance, and outbreak response.
Türkiye has implemented a national measles elimination programme since 2002, in parallel with the WHO European Region measles elimination goals. Despite substantial progress, measles elimination has not yet been achieved, and periodic resurgences have continued. National surveillance data show a marked peak in 2013, with cases continuing to be reported in subsequent years, while WHO/Europe data indicate another substantial resurgence during 2023–2024. Diyarbakır, a large province in southeastern Türkiye, has experienced measles activity in parallel with national epidemic patterns and may also be affected by additional outbreak dynamics related to population movement and migration from neighboring border countries [2, 3].
The differential diagnosis of patients presenting with fever and rash is broad, and establishing a definitive diagnosis based solely on clinical findings is often challenging [4]. When a suspected case is identified, rapid laboratory confirmation and timely implementation of control measures are critical to minimizing further transmission. However, in settings approaching measles elimination or where endemic transmission has been interrupted, laboratory confirmation of suspected cases becomes increasingly difficult. Therefore, the epidemiological context plays a crucial role in determining which cases warrant further testing [5].
Several factors have previously been associated with measles infection. Among these, vaccination status is the most frequently investigated, with being unvaccinated or incompletely vaccinated consistently identified as a major risk factor [6, 7]. In addition, other risk factors, such as contact with measles cases, travel history, hospital exposure, migrant status, social deprivation index, ethnicity, and exposure to fine particulate matter (PM2.5), have been reported in different studies [8–12]. Clinical features are most often examined by comparing vaccinated and unvaccinated individuals; unvaccinated individuals have been reported to exhibit symptoms more frequently and more severely [13, 14].
Infants represent a distinct age group requiring separate consideration because of waning maternally derived passive antibodies and variations across countries in the age at measles vaccination [15]. Studies focusing on infants have predominantly examined hospitalization and mortality outcomes, whereas investigations specifically addressing factors associated with laboratory-confirmed measles among suspected cases in this age group remain limited [16, 17]. A review of the literature reveals a notable paucity of studies that simultaneously examine clinical and epidemiological determinants of measles positivity and explore potential age-specific differences.
Recent global measles outbreaks further underscore the need for comprehensive, multifaceted assessments. The aim of this study was to identify age-specific epidemiological and clinical predictors of laboratory-confirmed measles among individuals who met suspected measles surveillance criteria, stratified into ≤ 1-year and > 1-year age groups. We anticipate that the findings will facilitate clinical decision-making for healthcare professionals and contribute to surveillance and targeted intervention strategies from a public health perspective.
Methods
Study design and population
This test-negative case–control study was conducted among individuals reported to the case-based measles/rubella surveillance system in Diyarbakır Province, Türkiye, from 2012 to 2024. Although the test-negative design is most commonly used in studies evaluating vaccine effectiveness, it has also been applied in investigations of factors associated with test positivity because it reduces bias related to healthcare-seeking behavior and provides more robust and valid estimates [18, 19].
In Türkiye, measles is monitored through a tailored laboratory-based surveillance system implemented within the framework of the World Health Organization’s Measles Elimination Program. Within this system, serological testing for measles and rubella is performed for individuals presenting to healthcare facilities with fever and maculopapular rash, or for those clinically suspected by a physician to have measles or rubella. Rubella IgM testing is performed systematically in parallel with measles IgM testing and is not restricted to measles-IgM-negative individuals. Serum samples in Diyarbakır are analyzed by enzyme-linked immunosorbent assay at the Public Health Laboratory affiliated with the Provincial Health Directorate.
Individuals residing in Diyarbakır who had serologic test results available were included in the study. Patients with positive measles-specific IgM antibody results were classified as case patients, whereas those with negative results were classified as control patients. Exclusion criteria were: (a) indeterminate or unavailable measles IgM antibody test results; (b) serological testing performed within 8–56 days after receipt of a measles-containing vaccine (MCV); and (c) measles IgM positivity with concurrent positive or indeterminate rubella IgM antibody test results. After exclusions, eligible individuals were stratified into two predefined age groups: ≤1 year and > 1 year. In each age stratum, all case patients were retained, and control patients were selected by systematic sampling at a 1:3 case-to-control ratio. Sampling was conducted across the entire 2012–2024 study period, rather than separately by calendar year or season. No individual matching was performed by calendar year, season, sex, district, or other variables. This resulted in 309 case patients and 927 control patients in the ≤ 1-year group, and 658 case patients and 1,974 controlpatients in the > 1-year group (Fig. 1).
Fig. 1.
Flowchart of sample selection in the measles test-negative case-control study
Data sources
Data collection tools included the measles/rubella surveillance database, the vaccination monitoring system, the laboratory information management system, and the standardized case investigation forms.
Implementation of measles-containing vaccine in Türkiye
Measles vaccination was first introduced in Türkiye in 1970, and both the number of doses and the timing of administration have changed over time. Until 1987, two doses were administered at 8 and 15 months of age. Between 1987 and 1998, a single dose was administered at 9 months of age. From 1998 to 2006, two doses were administered: one at 9 months and one during the first year of primary school. Since 2006, measles vaccination has been administered as part of the measles–mumps–rubella vaccine in two doses: at 12 months of age and during the first year of primary school. From 2020 through the end of the study period in 2024, the routine immunization schedule included the first MCV dose at 12 months and the second dose at 48 months. The 9–11-month measles-containing vaccine dose was recommended as an additional outbreak-control dose, independently of the routine MCV1–MCV2 schedule [20, 21]. Because Diyarbakır experienced periodic outbreaks and continued case reporting throughout 2012–2024, this additional dose was available and relevant in practice during the study period. Provincial or national supplementary immunization activities (SIAs) and catch-up campaigns were also implemented in different years. MCVs are provided free of charge in Türkiye.
Variables
Variables evaluated as potential predictors of laboratory-confirmed measles included age, sex, place of residence, immigrant status, vaccination status, clinical symptoms, hospitalization, season, and epidemic year. Age was calculated as the difference between the date of rash onset and the date of birth; the ≤ 1-year group was defined as individuals aged 0–12 months, with age expressed in months for this group and in years for individuals aged > 1 year. The place of residence was categorized as central or outlying districts within the province. Immigrant status was defined as individuals of foreign nationality present in the region, encompassing both foreign-born migrants and children born in Türkiye to immigrant parents. This demographic includes a diverse group, specifically individuals residing under official ‘temporary protection’ status, irregular migrants, and short-term foreign workers. However, in the specific context of the Diyarbakır province, the majority of this population consists of Syrian nationals living under temporary protection.
Vaccination status was categorized as unvaccinated, one dose at ≥ 12 months of age (MCV1), two doses at ≥ 12 months of age (MCV1 + 2), early dose (ED), early dose plus one dose (ED+MCV1), early dose plus two doses (ED+MCV1 + 2), and unknown. An early dose was defined as an MCV administered at 9–11 months of age.
An epidemic year was defined, based on provincial surveillance data, as a calendar year that included one or more months in which the incidence of laboratory-confirmed measles was ≥ 3 cases per 100,000 population [22]. Using this definition, 2013, 2015, 2023, and 2024 were classified as epidemic years during the study period; all other years were classified as non-epidemic years. The epidemic-year variable was assigned according to the year of rash onset. Clinical symptoms, categorized according to the suspected measles case definition used in surveillance, included fever; upper respiratory tract symptoms (URTS: coryza, cough, and/or conjunctivitis); and lymphadenopathy and/or arthralgia. Hospitalization was defined as admission at the initial clinical evaluation of individuals presenting with fever and maculopapular rash who met suspected measles/rubella surveillance criteria, before measles IgM results were available.
Statistical analysis
Descriptive statistics were summarized as frequencies and percentages for categorical variables and as medians (Q1–Q3) for non-normally distributed continuous variables. Comparisons between categorical variables were performed using Pearson’s chi-square test, Fisher’s exact test, or the chi-square test for trend, as appropriate. Continuous variables were compared using the Mann–Whitney U test.
Univariable and multivariable logistic regression analyses were conducted to identify clinical and epidemiological predictors of measles positivity. Two multivariable models were constructed: Model 1 included epidemiological variables, and Model 2 was an expanded predictive model that incorporated clinical variables. Variables were included in the multivariable models if they were statistically significant in univariable analyses (p < 0.05) or if they met all of the following criteria: an association with measles infection demonstrated in the literature, absence of a strong correlation with other variables (Spearman’s rho < 0.500), and balanced frequency distributions. Due to low cell counts, vaccination status categories ED and ED+MCV1 + 2 in the > 1-year age group, and the vaccination status in the ≤ 1-year age group, were excluded from multivariable analyses. Place of residence was excluded from univariable and multivariable logistic regression analyses in both age groups because of substantial missing data, while being presented descriptively and compared between case patients and control patients among individuals with available data. Season was dichotomized into spring/summer versus autumn/winter in regression analyses.
Model fit was assessed using the Hosmer–Lemeshow goodness-of-fit test, and the proportion of variance in measles positivity explained by the model was quantified using the Nagelkerke R² statistic. Crude odds ratios (cOR), adjusted odds ratios (aOR), and 95% confidence intervals (CIs) were reported. Statistical analyses were performed using SPSS version 22.0 (IBM Corporation, Armonk, NY, USA). A two-sided p-value < 0.05 was considered statistically significant.
Results
A total of 3,868 individuals were included in the study: 967 case patients and 2,901 control patients. Among the case patients, 32% (n = 309) were aged ≤ 1 year. The median age of case patients was 2.02 (0.8–6.8) years, ranging from 11 days to 49.7 years. Males accounted for 54.6% (n = 528) of case patients, and 9% (n = 87) were immigrants. Hospitalization was required for 12.5% (n = 121) of the case patients. The highest numbers of laboratory-confirmed measles cases were observed in 2013 (n = 491, 50.8%) and 2015 (n = 206, 21.3%). Seasonally, cases were most frequently identified during summer (n = 425, 44.0%) and spring (n = 400, 41.4%).
Results for the individuals aged ≤ 1 year
A comparison of the epidemiological and clinical characteristics of case patients and control patients aged ≤ 1 year is presented in Table 1. Among vaccinated infants, the median interval between vaccination and rash onset was 71 days (63–80). Among case patients in this age group, the proportion of hospitalization was 15.5% (n = 47).
Table 1.
Comparison of the epidemiological and clinical characteristics of case patients and control patients aged ≤ 1 year, Diyarbakır, 2012–2024
| Variable | Case patients (N:309) n (%)a |
Control patients (N:927) n (%)a |
p-valueb |
|---|---|---|---|
| Age [median (Q1–Q3)], months | 8.6 (7.2–10.3) | 7.5 (5.4–9.3) | < 0.001 |
| Age group, months |
< 0.001 p-trend < 0.001 |
||
| 0–6 | 71 (23) | 406 (43.8) | |
| 7–9 | 100 (32.3) | 249 (26.9) | |
| 10–12 | 138 (44.7) | 272 (29.3) | |
| Sex | 0.97 | ||
| Female | 136 (44) | 409 (44.1) | |
| Male | 173 (56) | 518 (55.9) | |
| Migrant status | 0.055 | ||
| Non-immigrant | 302 (97.7) | 920 (99.2) | |
| Immigrant | 7 (2.3) | 7 (0.8) | |
| Place of residencec | Valid: 75/309 | Valid: 218/927 | 0.03 |
| Central districts | 65 (86.7) | 163 (74.8) | |
| Outlying districts | 10 (13.3) | 55 (25.2) | |
| Epidemic year | < 0.001 | ||
| Yes | 270 (87.4) | 721 (77.8) | |
| No | 39 (12.6) | 206 (22.2) | |
| Season | < 0.001 | ||
| Autumn | 12 (3.9) | 80 (8.6) | |
| Winter | 33 (10.7) | 166 (17.9) | |
| Spring | 125 (40.4) | 250 (27) | |
| Summer | 139 (45) | 431 (46.5) | |
| Vaccination status | 0.001 | ||
| Unvaccinated | 308 (99.7) | 887 (95.7) | |
| Early dosed | 1 (0.3) | 40 (4.3) | |
| Clinical symptomsc | |||
| Fever | Valid: 281/309 | Valid: 785/927 | < 0.001 |
| Yes | 259 (92.2) | 623 (79.4) | |
| No | 22 (7.8) | 162 (20.6) | |
| Upper respiratory symptoms | Valid: 278/309 | Valid: 780/927 | < 0.001 |
| Yes | 238 (85.6) | 483 (61.9) | |
| No | 40 (14.4) | 297 (38.1) | |
| Lymphadenopathy/arthralgia | Valid: 149/309 | Valid: 524/927 | 0.02 |
| Yes | 9 (6) | 11 (2.1) | |
| No | 140 (94) | 513 (97.9) | |
| Hospitalizationc | Valid: 304/309 | Valid: 918/927 | < 0.001 |
| Yes | 47 (15.5) | 36 (3.9) | |
| No | 257 (84.5) | 882 (96.1) | |
aColumn percentages are presented
bP value for comparison between case patients versus control patients. Age compared using the Mann–Whitney U test; all other variables compared using Pearson’s chi-square test, Fisher’s exact test, or the chi-square test for trend
cFor variables with missing data, the number of individuals with available data out of the total number in each group is shown in the case patient and control patient columns. Percentages and p-values were calculated among individuals with available data
dEarly dose: Measles-containing vaccine administered at 9–11 months of age
In univariable logistic regression analysis, unvaccinated status was strongly associated with measles test positivity (cOR: 14.25; 95% CI: 1.95–104.05). In multivariable Model 1, which included epidemiological variables, significant predictors of measles positivity were ages 7–9 months (aOR: 2.18; 95% CI: 1.54–3.08) and 10–12 months (aOR: 2.86; 95% CI: 2.06–3.97); occurrence in an epidemic year (aOR: 1.92; 95% CI: 1.31–2.81); the spring–summer season (aOR: 1.95; 95% CI: 1.37–2.79); and immigrant status (aOR: 3.31; 95% CI: 1.09–10.02). In Model 2, which additionally included clinical variables, measles infection was not associated with immigrant status or epidemic year. In this model, older age and the spring–summer season remained significant predictors, while the presence of URTS (aOR: 3.20; 95% CI: 1.87–5.48) and hospitalization (aOR: 7.95; 95% CI: 4.21–15.00) were strong predictors of measles positivity (Table 2).
Table 2.
Predictors of laboratory-confirmed measles infection among suspected patients aged ≤ 1 year: univariable and multivariable logistic regression analysis, Diyarbakır, 2012–2024
| Univariable | Multivariable Model 1a |
Multivariable Model 2b |
|
|---|---|---|---|
| Variable | cOR (95%CI), p-value | aOR (95%CI), p value | aOR (95% CI), p value |
| Age group | |||
| 0–6 months | 1 | 1 | 1 |
| 7–9 months | 2.29 (1.63–3.23), < 0.001 | 2.18 (1.54–3.08), < 0.001 | 4.16 (2.33–7.41), < 0.001 |
| 10–12 months | 2.90 (2.09–4.01), < 0.001 | 2.86 (2.06–3.97), < 0.001 | 5.28 (3.03–9.20), < 0.001 |
| Migrant status | |||
| Non-immigrant | 1 | 1 | 1 |
| Immigrant | 3.05 (1.06–8.75), 0.039 | 3.31 (1.09–10.02), 0.034 | 3.72 (0.97–14.17), 0.054 |
| Epidemic year | |||
| No | 1 | 1 | 1 |
| Yes | 1.98 (1.36–2.86), < 0.001 | 1.92 (1.31–2.81), 0.001 | 1.15 (0.70–1.91), 0.57 |
| Season | |||
| Autumn–Winter | 1 | 1 | 1 |
| Spring–Summer | 2.12 (1.50–3.00), < 0.001 | 1.95 (1.37–2.79), < 0.001 | 2.08 (1.14–3.82), 0.017 |
| Vaccination statusc | NC | NC | |
| Early dosed | 1 | — | — |
| Unvaccinated | 14.25 (1.95–104.05), 0.009 | — | — |
| Symptoms | |||
| Fever | |||
|
No Yes |
1 3.06 (1.92–4.89), <0.001 |
1 0.79 (0.41–1.52), 0.49 |
|
| Upper respiratory symptoms | |||
|
No Yes |
1 3.66 (2.54–5.27), <0.001 |
1 3.20 (1.87–5.48), <0.001 |
|
| Lymphadenopathy/Arthralgia | |||
|
No Yes |
1 3.00 (1.22–7.38), 0.017 |
1 2.60 (0.89–7.63), 0.08 |
|
| Hospitalization | |||
| No | 1 | 1 | |
| Yes | 4.48 (2.84–7.07), < 0.001 | 7.95 (4.21–15.00),<0.001 |
Abbreviations: CI, confidence interval; cOR; crude OR; aOR, adjusted odds ratio; NC, not calculable
aModel 1: age group+migrant status+season+epidemic year (Hosmer–Lemeshow p:0.97; Nagelkerke R²:0.093)
bModel 2: Model 1 + symptoms+hospitalization (Hosmer–Lemeshow p:0.94; Nagelkerke R²:0.284)
cVaccination status was excluded from multivariable analysis due to highly imbalanced category distributions
dEarly dose: Measles-containing vaccine administered at 9–11 months of age
Results for individuals aged > 1 year
Comparisons of epidemiological and clinical characteristics between case patients and control patients aged > 1 year are shown in Table 3. Among case patients, the median interval between vaccination and rash onset was 1.5 (0.7–3.3) years for those who had received a single vaccine dose (n = 115 of 118 MCV1 case patients; vaccination-date data were missing for three individuals) and 3.7 (2.0–7.4) years for those who had received two doses (n = 4). The hospitalization rate among case patients in this age group was 11.4%.
Table 3.
Comparison of the epidemiological and clinical characteristics of case patients and control patients aged > 1 year, Diyarbakır, 2012–2024
| Variable | Case patients (N:658) n (%)a |
Control patients (N:1974) n (%)a |
p-valueb |
|---|---|---|---|
| Age [median (Q1–Q3)], years | 6.2 (3.6–10.3) | 3.2 (1.6–6.5) | < 0.001 |
| Age group, years |
< 0.001 p-trend < 0.001 |
||
| 1–4 | 188 (28.6) | 1132 (57.3) | |
| 5–9 | 272 (41.3) | 571 (29) | |
| 10–14 | 78 (11.9) | 181 (9.2) | |
| 15–19 | 60 (9.1) | 36 (1.8) | |
| 20–24 | 23 (3.5) | 20 (1) | |
| ≥ 25 | 37 (5.6) | 34 (1.7) | |
| Sex | 0.71 | ||
| Female | 308 (46.8) | 908 (46) | |
| Male | 350 (53.2) | 1066 (54) | |
| Migrant status | < 0.001 | ||
| Non-immigrant | 617 (93.8) | 1916 (97.1) | |
| Immigrant | 41 (6.2) | 58 (2.9) | |
| Place of residencec | Valid: 163/658 | Valid: 652/1974 | 0.53 |
| Central districts | 139 (85.3) | 543 (83.3) | |
| Outlying districts | 24 (14.7) | 109 (16.7) | |
| Epidemic year | < 0.001 | ||
| Yes | 606 (92.1) | 1091 (55.3) | |
| No | 52 (7.9) | 883 (44.7) | |
| Season | < 0.001 | ||
| Autumn | 24 (3.6) | 269 (13.6) | |
| Winter | 73 (11.1) | 458 (23.2) | |
| Spring | 275 (41.8) | 537 (27.2) | |
| Summer | 286 (43.5) | 710 (36) | |
| Vaccination statusd | < 0.001 | ||
| Unvaccinated | 169 (25.7) | 181 (9.2) | |
| MCV1 | 118 (17.9) | 1076 (54.5) | |
| MCV1 + 2 | 4 (0.6) | 126 (6.4) | |
| Early dose | 1 (0.2) | 33 (1.7) | |
| Early dose + MCV1 | 4 (0.6) | 193 (9.8) | |
| Early dose + MCV1 + 2 | 0 (0.0) | 14 (0.7) | |
| Unknown | 362 (55) | 351 (17.8) | |
| Clinical symptomsc | |||
| Fever | Valid: 608/658 | Valid: 1714/1974 | < 0.001 |
| Yes | 556 (91.4) | 1261 (73.6) | |
| No | 52 (8.6) | 453 (26.4) | |
| Upper respiratory symptoms | Valid: 602/658 | Valid: 1682/1974 | < 0.001 |
| Yes | 482 (80.1) | 816 (48.5) | |
| No | 120 (19.9) | 866 (51.5) | |
| Lymphadenopathy/arthralgia | Valid: 301/658 | Valid: 1201/1974 | 0.004 |
| Yes | 18 (6) | 32 (2.7) | |
| No | 283 (94) | 1169 (97.3) | |
| Hospitalizationc | Valid: 651/658 | Valid: 1928/1974 | < 0.001 |
| Yes | 74 (11.4) | 125 (6.5) | |
| No | 577 (88.6) | 1803 (93.5) | |
Abbreviations: MCV, Measles-containing vaccine
aColumn percentages are presented
bP value for comparison between case patients versus control patients. Age compared using the Mann–Whitney U test; all other variables compared using Pearson’s chi-square test, Fisher’s exact test, or the chi-square test for trend
cFor variables with missing data, the number of individuals with available data out of the total number in each group is shown in the case patient and control patient columns. Percentages and p-values were calculated among individuals with available data
dEarly dose: MCV administered at 9–11 months of age; MCV1: one dose at ≥ 12 months; MCV1 + 2: two doses at ≥ 12 months
In univariable analyses, each 1-year increase in age (cOR: 1.08; 95% CI: 1.07–1.10) and immigrant status (cOR: 2.19; 95% CI: 1.46–3.30) were associated with higher odds of measles positivity; however, both lost statistical significance in multivariable analyses. Vaccination status remained the strongest predictor of measles positivity across all analyses. Receipt of ED+MCV1 was not statistically associated with measles positivity compared with receipt of MCV1 + 2 in either univariable or multivariable analysis (univariable cOR: 0.65, 95% CI: 0.16–2.66; Model 2 aOR: 1.00, 95% CI: 0.21–4.70). This comparison included four measles-positive individuals in the ED+MCV1 group and four in the MCV1 + 2 group. In Model 2, unvaccinated individuals (aOR: 16.13; 95% CI: 4.77–54.54) and those with unknown vaccination status (aOR: 17.26; 95% CI: 5.14–57.88) demonstrated similarly strong associations with measles positivity (Table 4).
Table 4.
Predictors of laboratory-confirmed measles infection among suspected patients aged > 1 year: univariable and multivariable logistic regression analysis, Diyarbakır, 2012–2024
| Variable | Univariable | Multivariable Model 1a |
Multivariable Model 2b |
|---|---|---|---|
| cOR (95%CI), p-value | aOR (95%CI), p value | aOR (95% CI), p value | |
| *Age (years) | 1.08 (1.07–1.10), < 0.001 | 1.02 (0.99–1.04), 0.06 | 1.01 (0.99–1.04), 0.24 |
| Migrant status | |||
| Non-immigrant | 1 | 1 | 1 |
| Immigrant | 2.19 (1.46–3.30), <0.001 | 1.06 (0.65–1.74), 0.80 | 1.28 (0.72–2.27), 0.39 |
| Epidemic year | |||
| No | 1 | 1 | 1 |
| Yes | 9.43 (7.01–12.70), < 0.001 | 5.55 (4.05–7.62), < 0.001 | 4.06 (2.74–6.02), < 0.001 |
| Season | |||
| Autumn–Winter | 1 | 1 | 1 |
| Spring–Summer | 3.37 (2.67–4.26), < 0.001 | 2.75 (2.11–3.60), < 0.001 | 1.80 (1.21–2.69), 0.004 |
| Vaccination status c | |||
| MCV1 + 2 | 1 | 1 | 1 |
| MCV1 | 3.45 (1.25–9.51), 0.017 | 3.82 (1.37–10.67), 0.011 | 3.64 (1.09–12.15), 0.035 |
| Unvaccinated | 29.41 (10.63–81.34), < 0.001 | 23.71 (8.43–66.72), < 0.001 | 16.13 (4.77–54.54), < 0.001 |
| Early Dose+MCV1 | 0.65 (0.16–2.66), 0.55 | 0.73 (0.18–3.00), 0.66 | 1.00 (0.21–4.70), 0.99 |
| Unknown | 32.48 (11.87–88.85), < 0.001 | 23.24 (8.40–64.81), < 0.001 | 17.26 (5.14–57.88), < 0.001 |
| Symptoms | |||
|
Fever No Yes |
1 3.84 (2.83–5.20), < 0.001 |
1 1.89 (1.20–2.95), 0.005 |
|
|
Upper respiratory symptoms No Yes |
1 4.26 (3.41–5.32), < 0.001 |
1 1.34 (0.90–1.96), 0.13 |
|
|
Lymphadenopathy/ Arthralgia No Yes |
1 2.32 (1.28–4.20), 0.005 |
1 1.43 (0.68–3.02), p:0.34 |
|
|
Hospitalization No Yes |
1 1.85 (1.37–2.50), < 0.001 |
1 2.24 (1.39–3.62), 0.001 |
Abbreviations: CI, confidence interval; cOR; crude OR; aOR, adjusted odds ratio; MCV, Measles-containing vaccine
* Age was modeled as a continuous variable; ORs for age indicate the change in odds associated with each 1-year increase
aModel 1: age+migrant status+ vaccination status+season+epidemic year (Hosmer–Lemeshow p < 0.001; Nagelkerke R²:0.411)
bModel 2: Model 1 + symptoms+hospitalization (Hosmer–Lemeshow p:0.27; Nagelkerke R²:0.408)
cEarly dose: MCV administered at 9–11 months of age; MCV1: one dose at ≥ 12 months; MCV1 + 2: two doses at ≥ 12 months
Epidemic year and season remained consistent predictors of measles positivity across all models; in Model 2, epidemic year and season had aORs of 4.06 (95% CI 2.74–6.02) and 1.80 (95% CI 1.21–2.69), respectively. Although all evaluated clinical symptoms were associated with measles positivity in univariable analyses, fever was the predictor symptom in the multivariable analysis (aOR: 1.89; 95% CI: 1.20–2.95). Hospitalization, suggesting increased disease severity, remained independently associated with measles positivity (aOR: 2.24; 95% CI: 1.39–3.62) (Table 4).
Discussion
In this test-negative case–control study of individuals reported to the measles/rubella surveillance system as suspected cases, we identified age-specific clinical and epidemiological predictors of laboratory-confirmed measles infection. Among infants (≤ 1 year), the strongest predictors were age ≥ 7 months, the presence of upper respiratory tract symptoms, and hospitalization. In contrast, for individuals older than one year, vaccination status was the most prominent predictor, followed by epidemic year, fever, and hospitalization. Seasonality emerged as a consistent predictor across both age groups.
Infants accounted for approximately one-third of all measles cases in our study, which is concerning given their increased vulnerability to the potentially severe consequences of the disease [13, 15]. The decline in maternally derived antibodies by 6 months of age [23] likely contributed to the higher proportion of measles among infants aged 7–12 months. Although an outbreak-control early dose was available during the study period, documented early-dose vaccination was uncommon among infants in the surveillance dataset, which may also have contributed to susceptibility in this age group. In the > 1-year age group, measles cases were more likely to be older, consistent with prior observational studies [24, 25] and modeling studies assessing waning immunity [26]. Given that regional measles vaccination coverage increased from approximately 70% in 2005 to 85%, the higher proportion of measles cases observed among individuals aged ≥ 15 years may reflect historical immunity gaps.
As expected, our study found that measles was less frequently observed in individuals who had received at least one dose of an MCV. In the > 1-year age group, individuals with unknown vaccination status had measles positivity rates similar to those of unvaccinated individuals. This finding is consistent with a study by Wang et. in which vaccination status was reported as “none” (38%) or “unknown” (46%) for 84% of all reported measles cases between 2005 and 2014 [27]. Although unknown vaccination status may reflect true lack of vaccination or incomplete immunization in some cases, it may also result from incomplete documentation, migration, loss of vaccination records, limitations in registry linkage, or recall bias, particularly among older patients. Therefore, unknown vaccination status should not be interpreted as direct evidence of being unvaccinated or immunologically unprotected. Nevertheless, this group remains important for clinical assessment, surveillance, and catch-up vaccination decisions when reliable documentation of vaccination status is unavailable, particularly in exposure or other risk situations.
In Europe, measles remains frequently reported among immigrant populations, who often lack adequate immunological protection against the virus [28, 29]. In our study, immigrant status was more frequent among laboratory-confirmed cases, particularly among individuals aged > 1 year, and was associated with laboratory confirmation in univariable analyses; however, it was not identified as an independent predictor after adjustment. This finding should be interpreted cautiously and does not exclude the public health relevance of migration-related factors. Because of the test-negative design and the restriction of the study population to individuals presenting with maculopapular rash who met suspected measles surveillance criteria, our findings do not estimate population-level measles risk and therefore do not allow conclusions about the independent population-level risk attributable to immigrant status. Nevertheless, Diyarbakır has experienced migration from neighboring countries, particularly Syria, where WHO/UNICEF estimates indicate that MCV1 coverage remained below 80% throughout 2012–2023 [30]. Therefore, these subpopulations remain important from a public health perspective.
We assessed differences between central and outlying districts. The higher proportion of measles cases among infants in central districts may be attributed to high population density, cultural practices involving frequent home visits to newborns, and greater use of crowded tertiary care hospitals. This is supported by Zhang et al., who found that infants attending large municipal hospitals had a fivefold increased risk of measles [17]. Among older children, however, the main features associated with measles positivity among suspected cases-such as vaccination status and school-related exposures-may have been more uniformly distributed, potentially minimizing differences by residential location. Therefore, residing in a central or outlying district may no longer be a significant differentiating factor for laboratory confirmation among suspected cases in this age group.
Although measles typically peaks in late winter and spring in endemic temperate regions [31], we observed the highest case counts in spring and summer. Only a few studies have reported similar seasonal trends [4, 32]. These observed seasonal peaks are thought to result primarily from the influx of seasonal agricultural workers and increased travel to the region during these months. Another possible explanation, as reported by Yang et al., is that weather conditions influence the incidence of measles, with cases tending to occur immediately before or after cold-weather events [33]. Incorporating seasonality into transmission dynamics models may also inform the optimal timing of SIAs [34]. Therefore, additional research is needed to better understand the seasonal patterns and climatic mechanisms underlying measles transmission.
When clinical and epidemiological predictors were evaluated in combination, some age-specific differences in predictors of laboratory-confirmed measles among suspected cases emerged. Among infants presenting with a maculopapular rash, predictors included age 7–12 months, presence of upper respiratory tract symptoms, requirement for hospitalization, and presentation during peak measles seasons. For individuals older than one year, key predictors included unvaccinated or under-vaccinated status, fever, hospitalization, seasonality, and epidemic year. Among infants aged ≤ 1 year, univariable analysis showed that being unvaccinated was associated with 14-fold higher odds of measles positivity. Although vaccination status could not be included in multivariable modeling, univariable analysis suggests a potential association that warrants further investigation.
Extensive research has assessed the association between measles risk and immunization status, including the number of vaccine doses. In the study by Watabane et al., the odds of measles infection were 2.2 times higher among individuals who received a single dose and 5.8 times higher among those who were unvaccinated or whose status was unknown, compared with those who received two doses. Similarly, Mengistu et al. reported that unvaccinated individuals had approximately a 15-fold higher risk of measles than those who had received at least one dose [24, 35]. Other studies have reported that the odds of measles ranged from 2.6 to 17.2 among unvaccinated individuals [7, 36, 37]. Consistent with existing literature, our study found that, compared with individuals who received two doses, the odds of laboratory-confirmed measles were 3.6 times higher among those who received a single dose and approximately 17 times higher among unvaccinated individuals or those of unknown vaccination status. Although ED+MCV1 was not statistically associated with measles positivity compared with MCV1 + 2 in this surveillance dataset, this comparison was based on very few measles-positive individuals and produced imprecise odds-ratio estimates with wide confidence intervals. As the study was not designed or powered to compare the protection afforded by different vaccination schedules, and participants were restricted to suspected measles cases, this finding should not be interpreted as evidence of comparable protection or population-level vaccine effectiveness. It should be regarded only as the absence of a statistically significant difference in this dataset.
Differences in clinical predictors were also observed between infants and individuals older than one year of age. Among infants (≤ 1 year), upper respiratory tract symptoms were more strongly associated with laboratory-confirmed measles, whereas fever emerged as the more prominent symptom among individuals > 1 year. This may reflect modified clinical presentations, such as a shorter duration of fever in infants [38] or a lower frequency of respiratory symptoms in older age groups, following the introduction of vaccination [14]. In both age groups, however, hospitalization at initial presentation, reflecting clinical severity before laboratory results were available, was strongly associated with laboratory-confirmed measles among suspected cases.
Given the high basic reproduction number and the force of infection for measles, it is unsurprising that the epidemic year emerged as a significant predictor [39, 40]. Wallinga et al. further demonstrated that each additional susceptible individual beyond the epidemic threshold can result in nearly two additional infections [41]. However, in infants, the risk of infection can vary rapidly in just a few months—due to the waning of maternally derived antibodies and distinct caregiving or contact patterns—which may explain why epidemic year was a significant predictor only for the > 1-year age group. In contrast, seasonality may more sensitively reflect short-term shifts in transmission dynamics. Thus, among infants, seasonality may have been a stronger predictor of measles positivity than the epidemic year.
This study has several limitations. First, the extended observation period encompasses substantial shifts in vaccine coverage, SIAs, and migration patterns. Pooling data across heterogeneous epidemiological phases may have introduced temporal effect modification. Second, low cell counts in some vaccination subgroups led to the exclusion of certain categories, particularly infant vaccination status, from the multivariable analysis, which may have affected other estimates. Our study also has notable strengths. To our knowledge, this is the first study in Türkiye to assess a large sample and evaluate a wide range of predictors of laboratory-confirmed measles among suspected cases. Internationally, it is also among the few studies to have simultaneously assessed clinical and epidemiological characteristics stratified by age group.
Conclusion
Among individuals who meet suspected measles surveillance criteria and present with a maculopapular rash, evaluation of both clinical and epidemiological factors can support clinical decision-making, particularly in low-prevalence settings or where access to laboratory testing is limited. Our findings highlight age-specific differences in the predictors of laboratory-confirmed measles between infants and individuals older than one year; however, these results require further validation through additional studies. Identifying the most affected age groups and key epidemiological factors, such as seasonality and migration, may support the planning of catch-up campaigns and optimize the timing of SIAs at both national and regional levels.
Acknowledgements
Not applicable.
Abbreviations
- ED
Early dose
- MCV
Measles-containing vaccine
- SIAs
Supplementary immunization activities
- URTS
Upper respiratory tract symptoms
Author contributions
Conceptualization: D.Ç.; A.Y.Ç.; H.H.A.; M.C.; R.T.; Ö.F.Ş.; M.H.P. Data curation: Ç.D.A.; A.Y.Ç.; H.K.; R.T.; Ö.F.Ş.; M.H.P. Methodology: D.Ç.; H.H.A.; A.Y.Ç.; M.C. Formal analysis: D.Ç.; H.H.A.; Ç.D.A. Investigation: D.Ç.; H.K.; Ö.F.Ş.; M.H.P.; R.T. Supervision: H.H.A.; M.C. Writing the original draft: D.Ç.; H.K.; Writing, reviewing, and editing the final manuscript: All authors.
Funding
The authors received no financial support for the research or authorship.
Data availability
The datasets used and/or analysed during the current study are available from the corresponding author on reasonable request.
Declarations
Ethical approval
This study was conducted in accordance with the principles of the Declaration of Helsinki. Ethical approval for this study was obtained from the Ethics Committee of Diyarbakır Gazi Yaşargil Training and Research Hospital (approval date: 6 September 2024; protocol number: 154). Permission to conduct the study was also obtained from the Scientific Research Commission of the Diyarbakır Provincial Health Directorate (approval date: 8 November 2024; document number: E-61141975-799-259145179). Because this was a retrospective study using routine measles/rubella surveillance data, the requirement for informed consent to participate, including consent from parents or legal guardians for participants under 18 years of age, was waived by the Ethics Committee of Diyarbakır Gazi Yaşargil Training and Research Hospital.
Consent for publication
Not applicable. This manuscript does not contain any individual person’s identifiable data, images, videos, or case details.
Competing interests
The authors declare no competing interests.
Footnotes
Publisher’s note
Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.
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Associated Data
This section collects any data citations, data availability statements, or supplementary materials included in this article.
Data Availability Statement
The datasets used and/or analysed during the current study are available from the corresponding author on reasonable request.