Burnout in Colombian Health Professionals: A Psychometric and Descriptive Analysis

Abstract

This study examined burnout syndrome (BS) levels in 133 Colombian health professionals, aiming to identify its prevalence and evaluate the psychometric properties of the CESQT scale. The first aim was to determine the presence of BS during the initial peak of COVID-19 infections in Colombia, revealing generally low levels among participants. The second aim involved an exploratory factor analysis (EFA) of the CESQT, which confirmed its alignment with the theoretical and psychometric model of the original Spanish version and the Colombian adaptation, demonstrating good internal consistency and a four-dimensional structure. Finally, the relationship between the MBI-HSS and CESQT scales was explored, showing statistically significant correlations that support convergent validity. These findings enhance the understanding and measurement of BS in health professionals and offer insights for intervention, prevention, and mental health promotion aimed at improving healthcare services quality in Colombia.

1. Introduction

The World Health Organization (WHO) recently included burnout syndrome (BS) in the latest revision of the International Classification of Diseases (ICD-11, code QD85) [1]. Although BS it is not classified as a medical condition, it is defined as a psychological response to chronic occupational stress that has not been adequately managed [2]. This condition is particularly prevalent among professionals who provide services and are in contact with people, such as healthcare workers—who in Colombia are recognized as human talent in health (HTH). This group encompasses all personnel engaged in health promotion, education, prevention, diagnosis, treatment, rehabilitation, and palliation of the disease for all the inhabitants of the national territory within the organizational structure of the provision of health services [3].

One of the most widely used instruments to assess BS is the Maslach Burnout Inventory (MBI) [4], which has been adapted into several versions. The MBI-Human Services Survey (MBI-HSS) was designed for professionals of a wide range such as nursing staff, doctors, health assistants, social workers, health counselors, therapists, or other professions aimed at helping people. It evaluates three core dimensions: (1) emotional exhaustion, measuring feelings of being emotionally overloaded and exhausted by work; (2) depersonalization, assessing an insensitive and impersonal response towards the users of the service, care, or treatment. High scores on these two scales correspond to greater BS experienced; (3) personal accomplishment, measuring feelings of competence and successful achievements at work. Low scores on this scale correspond to higher BS.

Although the MBI remains a widely used instrument, it has some limitations. For instance, the depersonalization subscale often shows low internal consistency, as observed in previous validations in Colombia [5]. Concerns have also been raised regarding its factorial structure [6] and cross-cultural validity [7], particularly when adapted to Latin American populations. Moreover, the lack of universally accepted clinical cutoffs may limit its diagnostic utility in some contexts [8], and finally, its commercial license restricts broader access for public health research [9].

To address some of these limitations, Gil-Monte [10] developed the Burnout Syndrome Assessment Questionnaire (CESQT—Cuestionario para la Evaluación del Síndrome de Quemarse por el Trabajo), which assesses four dimensions: (1) illusion for work, defined as the desire to achieve work-related goals as a source of personal satisfaction. The items that make up this dimension are formulated in a positive sense, so that low scores indicate high levels of BS; (2) mental exhaustion, characterized by emotional and physical exhaustion due to working with people who present or generate problems, is an indicator of the decrease in personal affective resources; (3) indolence, referring to negative attitudes of insensitivity, indifference, and cynicism towards the people which the service is aimed at, where professionals show insensitive treatment towards clients and are not moved by the problems of others. High scores in these last two dimensions indicate BS; (4) guilt, defined as feelings of remorse, repentance for behaviors towards clients, and a feeling of having to apologize to those who carry out their work. This last dimension is not included when calculating the total level of BS; however, it allows establishing a critical profile of it.

For the above-mentioned reasons, it is urgent and necessary to have valid and reliable instruments to study the phenomenon during health crises in Colombian health professionals. Therefore, the present study has as its first aim to identify if the population studied presented BS during the initial peak of COVID-19 infections in the country. Likewise, it seeks to carry out an exploratory factor analysis (EFA) of the CESQT in Colombian health professionals to evaluate whether the construct behaves differently from the results found in a previous confirmatory factor analysis (CFA) [11]. Finally, it was explored whether there is a relationship between the constructs evaluated by the MBI-HSS and CESQT scales.

2. Materials and Methods

2.1. Design

The study incorporated a non-experimental, cross-sectional, quantitative, descriptive, and correlational design.

2.2. Participants

Given that the study was conducted during the initial peak of the COVID-19 pandemic, no a priori power analysis was performed to determine the required sample size. A non-probability and convenience sampling approach was adopted. Due to prevailing health conditions at the time, participant recruitment relied on voluntary responses to an online survey sent to over 6000 healthcare professionals via institutional mailing lists and social media. However, only 133 participants completed the questionnaire, resulting in a response rate of approximately 2.2%, highlighting the potential for self-selection bias [12], as individuals experiencing higher or lower BS may have had different motivations to respond. To minimize this bias, the questionnaire was anonymized, and it was clearly stated that participation was voluntary and unrelated to employment status.

Finally, analyses were based on data from 133 Colombian health professionals (94 from Antioquia, and 39 from other regions of the country). Of these, 82 were women (61.6%), aged 21 to 81 years (M = 43.7, SD = 12.1). Regarding occupation, 33.1% (n = 44) were anesthesiologists, 33.8% (n = 45) general practitioners, and 33.1% (n = 44) belonged to other professional categories (for more details of the sample see [13]).

2.3. Instruments

2.3.1. The Maslach Burnout Inventory-Human Services Survey (MBI-HSS)

Ref. [14] is made up of 22 items that measure the three dimensions of BS (emotional exhaustion, depersonalization, and personal accomplishment), which are individually rated using a Likert scale ranging from 0 (never) to 6 (always or every day). Cronbach’s alpha in the present study for the total scale was 0.86, showing good internal consistency overall.

2.3.2. Cuestionario Para la Evaluación del Síndrome de Quemarse por el Trabajo (CESQT)

Ref. [10], validated in surgical specialists in Antioquia, Colombia [11], contains 20 items that evaluate BS levels in four dimensions: illusion for work, mental exhaustion, indolence, and guilt, rated on a Likert-type scale of never (0), rarely (1), sometimes (2), frequently (3), and very frequently (4).

2.4. Procedure

After carrying out the pilot test, a questionnaire created with Google Forms^® was applied, during the months of September 2020 to February 2021. A survey link was disseminated via e-mail and social networks such as institutional websites, WhatsApp, Instagram, Facebook, etc. Additionally, participants were asked to voluntarily complete the questionnaire and encouraged to share it with other professionals in the country.

2.5. Statistical Analyses

There were no missing data on any of the questionnaires. Internal consistency (magnitude to which the items are correlated) was checked using Cronbach’s alpha for all scales. Descriptive statistics and normality tests (Shapiro–Wilk) were conducted to identify the type of distribution of each variable finding that they do not present a normal distribution for which non-parametric statistics were performed. The structure of the correlation matrix of all the variables was verified through Spearman correlations. An exploratory factor analysis (EFA) of the CESQT was performed using a principal component analysis extraction method and orthogonal rotation (varimax). Although it is recognized that the underlying dimensions of SB could be theoretically interrelated and that in that case the use of an oblique rotation would be more convenient, this decision was made to facilitate the interpretation of the factors, assuming uncorrelated components, as is usual in initial psychometric explorations [15]. The adequacy of the EFA was verified using the sample adequacy criteria with the Kaiser–Meier–Olkin (KMO) coefficient and Bartlett’s test of sphericity. For all analyses, p values < 0.05 were considered statistically significant. Analysis was performed using STATA 16 software.

3. Results

3.1. Descriptive Statistics

Table 1. Descriptive statistics and internal consistency of the MBI-HSS and CESQT.

Subscales	Minimum	Maximum	Mean	Standard Deviation	Cronbach’s Alpha
Emotional Exhaustion	0	54	17.37	12.65	0.90
Depersonalization	0	30	3.32	4.34	0.50
Personal accomplishment	0	48	37.93	9.66	0.87
Illusion for work	0	20	15.43	4.50	0.93
Mental exhaustion	0	16	6.74	4.67	0.91
Indolence	0	24	5.07	3.97	0.82
Guilt	0	20	3.63	3.52	0.87
Burnout syndrome	0	58	27.26	7.38	0.89 a

^a Cronbach’s alpha for the total CESQT (BS) does not include the guilt subscale.

presents descriptive statistics for each scale. In the MBI-HSS, for emotional exhaustion, low levels (scores of 0–18) were reported by 78 participants (58.6%), medium levels (scores of 19–26) by 22 participants (16.5%), and high levels (scores of 27–54) by 33 participants (24.8%). Regarding depersonalization, 102 participants (76.7%) reported low levels (scores of 0–5), 18 participants (13.5%) reported medium levels (scores of 6–9), and 13 participants (9.8%) reported high levels (scores of 10–30). Finally, for personal accomplishment, low levels (scores of 0–33) were reported by 34 participants (25.6%), medium levels (scores of 34–39) by 24 participants (18.0%), and high levels (scores of 40–56) by 75 participants (56.4%). Considering the cut-off points and reference values established by the creators of the inventory, in general the present study found low emotional exhaustion and depersonalization levels and high personal accomplishment levels. Cronbach’s alpha values indicated good internal consistency for all subscales except depersonalization.

For the CESQT, participants reported moderately high illusion for work, medium BS, and low scores on other subscales. Based on CESQT cut-off points, only three people (2.26%) were at a very low level (scores < 11); 108 (81.2%) at a low level (scores of 11–33); 22 (16.54%) at an intermediate level (scores of 34–66); and no one presented high scores (67 to 89) or critical scores (>89). In addition, Cronbach’s alpha values showed good internal consistency in all scales.

3.2. Exploratory Factor Analysis

The objective of a factor analysis is to look for patterns of association of variables or items to reduce their information to underlying factors (principal components), which are a linear combination of the initial items. That is, it is to maximize the amount of variance explained by the minimum number of factors (parsimony) [16]. Following the recommendations of García Borrero et al. [11] (p. 8), an AFE of the CESQT was performed to determine whether the construct behaves differently from the results found in their CFA; the results will be discussed in Section 4 along with the procedures and conceptual decisions that have been followed.

The first step was to verify whether the data structure is suitable for factor analysis, for which sample adequacy measures such as Bartlett’s sphericity test, and the Kaiser–Meyer–Olkin (KMO) adequacy test was used. The first test assumes that if there is statistical significance the variables are different, and in the present study a value of p < 0.0001 was obtained. For the second test, which indicates the degree of correlation of a variable with the rest of the variables, Kaiser [17] recommends considering the appropriate matrix to carry out the factorization when the value of this indicator is greater than or equal to 0.80, and in the present study, all the items showed higher values and the measure of total adequacy was 0.88.

Following this, the analysis of the correlation structure was carried out by extracting factors with a principal component method, which establishes uncorrelated linear combinations of the observed variables. The first component has the maximum variance and the successive ones progressively explain smaller proportions of the variance and are not correlated with each other [18]. The criterion for selecting the number of factors was the Kaiser–Guttman rule, according to which the threshold for factor extraction is usually set at a minimum of 60% [19] or 70–80% [20]; in this study only the first four factors with eigenvalues greater than or equal to 1 were taken, which explain 70.2% of the total variance [21] (

Table 2. Variance explained by the CESQT’s items.

Factor	Eigenvalue	Difference	Proportion	Cumulative
1	7.56541	4.20572	0.3783	0.3783
2	3.35969	1.39253	0.1680	0.5463
3	1.96717	0.81603	0.0984	0.6446
4	1.15114	0.29752	0.0576	0.7022
5	0.85362	0.17534	0.0427	0.7449
6	0.67828	0.04542	0.0339	0.7788
7	0.63285	0.10069	0.0316	0.8104
8	0.53217	0.03923	0.0266	0.8370
9	0.49294	0.07881	0.0246	0.8617
10	0.41412	0.00507	0.0207	0.8824
11	0.40905	0.07181	0.0205	0.9028
12	0.33724	0.05935	0.0169	0.9197
13	0.27789	0.01227	0.0139	0.9336
14	0.26562	0.01300	0.0133	0.9469
15	0.25262	0.04548	0.0126	0.9595
16	0.20713	0.03106	0.0104	0.9698
17	0.17608	0.01104	0.0088	0.9787
18	0.16504	0.00533	0.0083	0.9869
19	0.15971	0.05747	0.0080	0.9949
20	0.10224		0.0051	1.0000

Extraction method: LR test independent vs. saturated. Chi2 (190) = 1802.37 Prob > chi2 = 0.0000.

).

The factor loadings are equivalent to communality, which is defined as the percentage of variability of each variable or item that can be explained by the extracted factors [16]. To be able to interpret the correlation matrix between each of the new factors obtained and the original variables, a varimax orthogonal rotation method was used, which assumes that the factors are independent from each other, from a conceptual point of view. On the other hand, Field [22] recommends interpreting only those factor loads with an absolute value greater than 0.40, or that account for 16% of the variance of the variable, while correlations equal to 0.45 represent poor variability (20%), correlations of 0.55 good variability (30%), correlations of 0.63 very good variability (40%), and correlations of 0.71 optimal variability (50%). Then, it can be concluded that a factor that contains at least three items with loads greater than 0.60 are enough to assume them as good indicators of the construct and its validity [13]. As can be seen in

Table 3. CESQT exploratory factor analysis.

Items	Factor 1 Illusion for Work	Factor 2 Guilt	Factor 3 Mental Exhaustion	Factor 4 Indolence
1. Mi trabajo me supone un reto estimulante	0.85	−0.03	−0.13	−0.01
2. No me dan ganas de atender a algunas personas en mi trabajo	−0.11	0.24	0.37	0.43
3. Creo que muchas de las personas a las que atiendo en el trabajo son insoportables	−0.14	0.30	0.11	0.76
4. Me preocupa el trato que he dado a algunas personas en el trabajo	−0.15	0.56	0.24	0.47
5. Veo mi trabajo como una fuente de realización personal	0.89	0.01	−0.00	−0.06
6. Creo que los familiares de las personas a las que atiendo en el trabajo son unos pesados	0.01	0.16	0.21	0.78
7. Pienso que trato con indiferencia a algunas personas a las que atiendo en mi trabajo	−0.18	0.52	0.15	0.45
8. Pienso que estoy saturado(a) por el trabajo	−0.12	0.07	0.89	0.16
9. Me siento culpable por alguna de mis actitudes en el trabajo	−0.19	0.74	0.23	0.19
10. Pienso que mi trabajo me aporta cosas positivas	0.88	−0.24	−0.04	−0.09
11. Me dan ganas de ser irónico(a) con algunas personas que atiendo en el trabajo	−0.18	0.21	0.23	0.76
12. Me siento agobiado(a) por el trabajo	−0.12	0.17	0.78	0.16
13. Tengo remordimientos por algunos de mis comportamientos en el trabajo	−0.10	0.78	0.16	0.34
14. Etiqueto o clasifico a las personas a las que atiendo en el trabajo según su comportamiento	−0.02	0.47	0.13	0.45
15. Mi trabajo me resulta gratificante	0.93	−0.02	−0.15	−0.08
16. Pienso que debería pedir disculpas a alguien por mi comportamiento	0.01	0.84	0.08	0.09
17. Me siento cansado(a) físicamente en el trabajo	−0.14	0.19	0.87	0.06
18. Me siento desgastado(a) emocionalmente	−0.12	0.20	0.84	0.18
19. Me siento ilusionado(a) por mi trabajo	0.80	0.00	−0.26	−0.09
20. Me siento mal por algunas cosas que he dicho en el trabajo	0.03	0.71	0.32	0.15

in this study, most items present values above 0.7, and only the items 2, 7, and 14 present poor variability with respect to factor 4.

The last step of the EFA is the naming of the factors, and according to the dimensions proposed by the creators of the CESQT [23], the authors of the used version [11], and the results of the present study, the composition of the scales or dimensions with their respective items would be the following: factor 1 is named illusion for work (items 1, 5, 10, 15, and 19), factor 2 guilt (items 4, 9, 13, 16, and 20), factor 3 mental exhaustion (items 8, 12, 17, and 18), and factor 4 indolence (items 2, 3, 6, 7, 11, and 14).

3.3. Correlations

As shown in

Table 4. Spearman correlations between the MBI-HSS and CESQT scales.

	EE	D	PA	IW	ME	I	G
D	0.47 **
PA	−0.32 *	−0.23 *
IW	−0.42 **	−0.24 *	0.60 **
ME	0.79 **	0.36 **	−0.34 **	−0.36 **
I	0.52 **	0.56 **	−0.36 **	−0.31 *	0.48 **
G	0.43 **	0.37 **	−0.28 *	−0.32 **	0.44 **	0.64 **
BS	0.69 **	0.41 **	−0.13	0.07	0.78 **	0.67 **	0.49 **

** p < 0.0001. * p < 0.05. EE = emotional exhaustion; D = depersonalization; PA = personal accomplishment; IW = illusion for work; ME = mental exhaustion; I = indolence; G = guilt; BS = burnout syndrome.

, the matrix of correlation coefficients between the variables confirms that the application of a factorial analysis is appropriate (<0.30 and >0.80), in addition to verifying that there is no co-linearity between the scales. Likewise, statistically significant relationships show convergent validity between the constructs that both instruments measure. Specifically, it can be observed that the scales of personal accomplishment and illusion for work presented a moderate positive correlation with each other, and negative correlations with the rest of the dimensions, as would be expected theoretically. However, they were not significantly related to BS. On the other hand, most of the statistically significant and positive correlations between the other subscales had moderate association coefficients, while only the correlations between mental exhaustion with emotional exhaustion and BS were strong.

4. Discussion

The first aim of the present study was to determine whether Colombian health professionals experienced BS during the initial peak of the COVID-19 pandemic, for which it was found that in the MBI-HSS more than half presented low emotional exhaustion and depersonalization and high personal accomplishment. As in the CESQT, there were medium-high scores in illusion for work and low scores in BS and the other subscales. As reported by Gil-Monte et al. [23], exposure to work conditions characterized by a lack of clarity, decision-making, autonomy, and social support, in addition to high pressure and demand, can generate moderate levels of BS, but not necessarily high or critical, as observed in the present study, with low scores in 83.5% and medium scores in 16.5% of the participants, ranking below the prevalence previously reported in the country, which is between 17.6% and 45% [24,25].

Therefore, it can be concluded that the Colombian health professionals evaluated in the present study did not present BS, since theoretically low scores would be expected in illusion for work and personal accomplishment, and high scores in the rest of the dimensions [26]. Likewise, the high scores in illusion for work and personal accomplishment could be explained by the fact that the Colombian health professionals could have positive expectations that allow them to anticipate job satisfaction, as has been reported in other studies [27,28], where despite the fact that the health crisis could generate discomfort and uncertainty, there were also positive feelings derived, for example, from working conditions, the fact of maintaining employment and economic stability, etc. However, another possible explanation may be that in professional contexts, participants may respond in ways that are perceived as socially acceptable or desirable, especially when items address ethical behavior or negative emotions [29]. In addition, regarding the contextual conditions under which the present study was conducted, workers may have minimized their discomfort out of a sense of duty or fear of institutional judgment.

The second aim of this study was to perform an EFA of the CESQT in Colombian health professionals, finding similar results to the CFAs of Gil-Monte et al. [23] and García Borrero et al. [11], where four dimensions corresponding to those of the theoretical model that precedes the psychometric model were verified. However, as observed in the present study, regarding the fact that items 2, 7, and 14 presented poor variability with respect to factor 4 (indolence) and factor 2 (guilt), from a statistical point of view, it can be explained since in simple, clear, and interpretable multidimensional solutions it would be desirable for each item to be mainly a good indicator of a single factor. However, items generally also have secondary weights on other factors. The salient (highest) weight reports on the factor that this item primarily assesses, while the secondary weights (i.e., not as high as the salient, but still high enough to be meaningfully interpreted) reflect the influence of other factors and, therefore, also provide substantial information to estimate them. Likewise, this can lead to little variance in the responses since if the majority responds similarly (e.g., everyone marks “never”), the item does not discriminate well, and therefore its factor loading will be low [30].

Other possible reasons for items 2, 4, and 7 showing weak loadings on factors 2 and 4, corresponding to guilt and indolence, respectively, may be due to (a) semantic ambiguity in the wording or double emotional interpretation that could have confused respondents. For example, item 7 (“I think I treat some people with indifference…”) may overlap with feelings of guilt (factor 2) and with indolence (factor 4), generating cross-sectional or weak loadings; (b) overlap between dimensions, since if an item reflects aspects in more than one factor, it may not be strongly associated with just one, reducing its factor loading; and that (c) some expressions may not have the same cultural resonance or may not be considered appropriate in certain contexts, affecting the way respondents respond [30].

From a theoretical point of view, the CESQT model developed by Gil-Monte et al. [23], proposes that in the face of high job stress, the first responses correspond to cognitive deterioration (low scores in illusion for work) and affective deterioration (high scores in mental exhaustion). Secondly, negative attitudes towards people or work (high levels of indolence), and later, the appearance of feelings of guilt, could appear in some people (although not in all), with which it is possible to identify two profiles in the BS development process, one that does not include feelings of guilt and the other that includes them. Moreover, it is in the latter where those participants with high scores in both the indolence factor and the guilt factor reported in this study would be found, as found in the study by Gil-Monte et al. [23], who also found a high correlation between both dimensions.

In addition, the low internal consistency of the BMI-HSS depersonalization subscale reported in the literature ([11] (p. 7) and [5]), and found in the present study (α = 0.50), could be explained due to the fact that it contains only five items, some of which are ambiguous or poorly culturally adapted, and as stated by Gil-Monte et al. [23] when they argue that although this is one of the most widely used instruments to assess BS, it presents some psychometric insufficiencies when it is adapted to other languages, which may be occurring in the version used in the present study. In addition, added to other aspects of the participants, since the depersonalization dimension, which theoretically refers to the degree to which each person recognizes attitudes of coldness and distance towards others and/or their work, in this case it could be a defensive mechanism against the emotional burden caused by the severity of the COVID-19 pandemic, patient deaths, fear of contagion, etc. Another hypothetical explanation would be that, taking into account the proposal by Gil-Monte et al. [23], by including the guilt dimension in the BS profile, this could overlap when interpreting responses of coldness, not from an objective point of view of said behavior, but from a moral vision that makes the professional feel remorse, as may be occurring with items 7 and 14 of the CESQT, which present factor loads in both factor 4 (indolence) and factor 2 (guilt).

The last aim of this study was to explore whether there is a relationship between the constructs evaluated by the MBI-HSS and CESQT; as Gil-Monte et al. [23] stated in terms of content, that there is a similarity between both instruments, especially between mental exhaustion with emotional exhaustion, indolence with depersonalization, and between illusion for work and personal accomplishment, which were confirmed from the statistically significant correlations reported in Table 4. However, the fact that personal accomplishment and illusion for work were not significantly related to BS, although contrary to expectations, is not entirely inconsistent, since as previously stated [31], it is the dimensions with negative content that are considered basic components of BS, and not necessarily the low scores in the positive dimensions. On the other hand, it is noteworthy that the correlations between mental exhaustion with emotional exhaustion and BS were strong, in contrast to the results of García Borrero et al. [11], who found high correlations between mental exhaustion and guilt with indolence.

Limitations and Future Research

One of the main aspects to highlight in this study was that the variables were assessed during a precise moment of the global health crisis, namely the initial peak of infections in the COVID-19 pandemic, in different professions of Colombian health professionals. However, most of the participants were women, anesthesiologists, and physicians, from Antioquia. In addition, the small sample size and the type of non-probabilistic sampling prevent the generalization of results, which is why it is recommended to expand the sample not only in the number and sociodemographic characteristics of the participants, but also to extend the study to other regions of the country and other professions.

Another relevant aspect to interpret the results of this study with caution is that, although the sample adequacy tests to perform the EFA yielded adequate results, the sample size was less than 200 participants, and therefore, for future studies the sample size could be expanded and a CFA could be performed to correct the deficiencies of the EFA. Likewise, the use of a varimax orthogonal rotation would have to be reviewed in depth, because, although the items are part of the same instrument, with which they would be correlated with each other, theoretically they measure different aspects of BS, so different rotation types should be tested and all their values reported.

On the other hand, the fact that higher scores were presented in the positive dimensions of BS rather than in the negative ones could be due to a bias caused by the type of convenience sampling and social desirability, since the situation of job uncertainty at the level of hiring and other novel conditions from the COVID-19 health crisis could have led the participants in this study to respond thinking that their bosses would know the answers (despite the fact that the questionnaire did not require identification data and was answered individually) and therefore not give completely honest answers.

Lastly, it would be interesting to carry out longitudinal studies to verify BS, in addition to including other variables and comparative analyses by gender, age, region, profession, etc., that make it possible to identify if BS occurred specifically due to the pandemic or if there were other risk and protective factors that influenced the response to the global health crisis.

5. Conclusions

The fact that low levels of BS have been reported in Colombian health professionals during the COVID-19 pandemic would be accounting for the existence of protective factors [31], possibly evidenced in the high scores in the illusion for work and personal accomplishment scales. Despite the previously mentioned limitations, it was possible to verify that the CESQT in Colombian health professionals presents similar psychometric properties to the original version created with the Spanish population and the version validated in surgical specialists from the department of Antioquia, Colombia, since it presents a good internal consistency, the underlying structure of four dimensions was confirmed, which are significantly related to the MBI-HSS scales, and therefore it is a valid and reliable instrument to evaluate BS in different occupational and sociocultural groups.

The results of this study, in addition to confirming a previous theoretical and psychometric model, serve at a practical level to have a valid and reliable BS measurement instrument, beyond simply having a version in Spanish, as well as one that also provides elements for the formulation of plans for intervention, prevention, and promotion of mental health, which in turn could impact the well-being and the quality of the services provided by the health professionals in the country. Regarding public occupational health policies and clinical management in healthcare institutions, the validation of the CESQT could support its use not only in research but also as a tool for routine psychological screening and early detection of burnout syndrome in clinical settings, to inform workforce planning, prevent professional attrition, and guide investments in workplace well-being, especially in rural areas or those with limited healthcare system resources.

At the institutional level, it would allow for periodic assessment of psychosocial risk among healthcare workers, particularly in high-stress settings such as emergency departments, intensive care units, or during public health emergencies. By identifying individuals in the early stages of burnout—especially those with low scores on “work enthusiasm” and high scores on “mental exhaustion” or “indolence”—interventions can be designed targeting the most affected areas. Furthermore, the scale’s ability to detect nuanced profiles, including the guilt-related dimension, allows professionals to tailor support interventions to the cognitive-affective and behavioral aspects of burnout syndrome. This is particularly relevant for the development of mental health programs, psychoeducational workshops, or individualized referral systems to occupational psychologists. Furthermore, longitudinal administration of the CESQT could be integrated into human resources protocols (e.g., during onboarding and annual evaluations) to monitor changes over time and assess the impact of organizational changes, workload, or wellness programs on staff well-being.