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Smiling face as anchor for pain intensity scales
Letters to the Editor
The suggestion for control groups such as other chronic pain patients and psychological distressed patients is well appreciated and might be an advantage in future studies. Nevertheless, it is extremely important to match the groups so that they are equally distressed and to make sure that the physiology of pain in the control group cannot affect the response to experimental stimuli per see. For instances the responsiveness of dorsal horn neurons might be enhanced in both the group of neck pain patients and low back pain patients in the study of Carragee et al. (1998) and then affect the response to the experimental stimuli in both groups. Using the suggested approach with chronic pain patients as controls we end up with another problem: How to show that the physiology of pain does not corrupt the control group? This can be compared to the problem raised in the comments: How to show that the psychological factors do not corrupt the control group? Hopefully our study and the study by Carragee et al. (1998) together will inspire to new studies on central sensitization in chronic musculoskeletal pain patients. References Carragee EJ. False positive lumbar discography in selected patients without low back pain complaints, San Francisco, CA: North American Spine Proceedings, Annual Meeting, 1998. Graven-Nielsen T, Kendall SA, Henriksson KG, Bengtsson M, SoÈrensen J, Johnson A, Gerdle B, Arendt-Nielsen L. Ketamine reduces muscle pain, temporal summation, and referred pain in ®bromyalgia patients. Pain 2000;85:483±491. Hoheisel U, Mense S, Simons DG, Yu X-M. Appearance of new receptive ®elds in rat dorsal horn neurons following noxious stimulation of skeletal muscle: a model for referral of muscle pain? Neurosci Lett 1993;153:9±12. Johansen MK, Graven-Nielsen T, Olesen AS, Arendt-Nielsen L. Generalised muscular hyperalgesia in chronic whiplash syndrome. Pain 1999;83:229±234. Sorensen J, Graven-Nielsen T, Henriksson KG, Bengtsson M, Arendt-Nielsen L. Hyperexcitability and ®bromyalgia. J Rheumatol 1998;25:152± 155.
Thomas Graven-Nielsen a, Mona Koelbaek Johansen b*, a Anders S. Olesen c, Lars Arendt-Nielsen a
Center for Sensory-Motor Interaction, Laboratory for Experimental Pain Research, Aalborg University, Denmark b Department of Rheumatology, Aalborg Hospital, Denmark c The Pain Clinic, Aalborg Hospital, Denmark
* Corresponding author. PII: S 0304-395 9(00)00374-2
295
PAIN 4321
Smiling face as anchor for pain intensity scales
One of the most important challenges facing pediatric health care professionals is the identi®cation of a valid and easy-to-use scale to assess pain in children. We are concerned that the research by Chambers and others (1999) adds to this challenge by asserting largely unsubstantiated theoretical assumptions for their directional hypothesis and questionable statistical ®ndings to support their conclusions. The authors assert that `faces scales beginning with a smiling no pain face confound non-noxious affective states with pain; for example, children who are not in pain are not necessarily happy.' They further assume that a neutral `no pain' face avoids introducing an affective component to the pain scale. However, there is no evidence presented in this study or in their previous study (Chambers and Craig, 1998) that children view a `neutral face' as devoid of emotion. In fact, one can argue that children may attribute emotions such as fear, anger, sadness, or unhappiness to a facial expression that adults describe as neutral, thus confounding the assessment of pain with negative emotions. The authors also predicted the direction of their hypothesis: `scales with smiling `no pain' faces would elicit higher pain ratings, in comparison to ratings using scales with neutral `no pain' faces.' However, any ®ndings about the scores from the ®ve scales can only be compared to each other. Therefore, it is just as tenable to conclude that the neutral face scales produced falsely low pain scores as it is to conclude that the smiling face scales produced falsely high pain scores. Unless the scores were compared to a `gold standard' instrument for assessing pain, it is premature to conclude in which direction the initial `anchor' face biases a child's rating of pain intensity. Following the authors' logic that `happiness' re¯ected in a smiling face causes children to choose facial expressions toward the more distressed end of the scale, it is also possible that socalled neutral faces may less clearly represent `no pain' to children who tend to choose faces closer to the less distressed end of the scale. Although the authors reported an unsubstantiated signi®cant main effect for the differences between the two smiling scales and the three neutral scales using the Student± Newman±Keuls post-hoc test, we question these results. First of all, we cannot visualize how the data (see Table 1) with such small differences among the standardized means (maximum difference of 0.54) but with relatively large and consistent standardized standard deviations for each scale (range of 2.99±3.33) could yield signi®cant results for the Student±Newman±Keuls test. Secondly, the authors gave only verbal statements concerning the results of the post-hoc test without any numerical support. Therefore, we subjected the data provided in Table 1 to ®ve multi-
296
Letters to the Editor
ple comparison tests: the Fisher's Least Signi®cant Difference test, the Duncan's New Multiple Range test, the Student±Newman±Keuls test, the Tukey's Honestly Significant Difference test, and the Scheffe's test to test for significant differences among the scales. These tests are listed in ascending order of their degree of conservatism (Dowdy and Wearden, 1991). Prior to performing these tests, the pain scales were arranged in ascending order of the size of their means, such that the Lebaron and Zeltzer scale became scale 1 and the Wong and Baker scale became scale 5. The difference between the mean of scale 5 and scale 1 was 0.54, which represents the maximum difference between any two scales. For the Fisher, Tukey, and Scheffe post-hoc tests, the size of the con®dence interval is constant for all pairs of scales and depends only on the test involved. If neither the Fisher test (most liberal) nor the Scheffe test (most conservative) showed any of the differences to be signi®cant, then the obvious conclusion would be that the authors' statement of a signi®cant Student±Newman± Keuls test is not supported. The computed con®dence interval for the difference in means for scale 5 and scale 1 were 0.8488, 1.1271, 1.4077, 1.4077, and 1.6641 for the ®ve multiple comparison tests, respectively. It should be noted that when comparing the difference in means between scale 5 and scale 1 all the post-hoc tests yielded critical intervals larger than the maximum difference of 0.54. The Student±Newman±Kuels procedure yields different critical intervals depending on the span of the two ranked averages being compared. In this case, the critical intervals for two ranked scales spanning from one to four ranks were 1.0114, 1.2092, 1.3256, and 1.4077, while the
maximum difference in means for any two scales with corresponding rank span were 0.4700, 0.5100, 0.5200, and 0.5400, respectively. It can be seen that there is no signi®cant difference in means. This ®nding brings up the question of whether the original ANOVA was carried out correctly since it indicated a signi®cant main effect for difference between the scales. However, the reported F 3:27; 225:45 5:14, P 0:001 calculation differs from the values originally reported in a prepublication paper we read from the authors of F 4; 276 14:35, P , 0:001. When we conveyed our initial concern about the Student± Newman±Keuls post-hoc test to the authors, they replied that their calculation had been re-analyzed by several statistical consultants and that they `remained con®dent that our analyses are appropriate and the results accurate.' However, not only did their ®ndings change, they also expanded. An addition to their published paper is their ®nding of a significant main effect for gender, F 1; 69 3:70, P 0:05. However, this is incorrect since the critical F(1,69) at alpha 0.05 is 3.980. In summary, our post-hoc testing of the data does not support the authors' conclusion, and the discrepancies between the prepublished and published data need clari®cation. We have recently completed a metanalysis of 30 published and 19 unpublished studies using the WongBaker FACES scale to measure pain in children and adults. Although none of these studies compared the Wong-Baker FACES to the same four facial scales in this study, the research included comparisons to most well-established scales, such as numeric, verbal-description, visual-analogue, color, photographic (Oucher), and poker-chip scales.
Table 1 Descriptive statistics of children's and parents' ratings for each of the ®ve faces scales Bieri et al. (0-6)
Wong and Baker (0-5)
Children's ratings Mean (raw) Mean (standardized/10) a Median (raw) Median (/10) Standard deviation (/10) Skew Kurtosis
1.49 2.49* 1.00 1.67 3.15 1.30 0.68
1.51 3.01** 1.00 2.00 3.18 0.99 0.10
Parents' ratings Mean (raw) Mean (standardized/10) a Median (raw) Median (/10) Standard deviation (/10) Skew Kurtosis
2.09 3.49* 2.00 3.33 3.04 0.83 20.31
2.15 4.29** 2.00 4.00 2.82 0.50 20.38
Maunuksela et al. (0-4)
LeBaron and Zeltzer (0-4)
Kuttner and LePage (0-4)
1.19 2.97** 1.00 2.50 3.33 0.85 2 0.45
0.99 2.47* 1.00 2.50 3.14 1.33 0.79
1.00 2.50* 1.00 2.50 2.99 1.26 0.89
1.47 3.67* 1.00 2.50 2.97 0.90 20.01
1.21 3.03*** 1.00 2.50 2.89 1.09 0.61
1.24 3.10*** 1.00 2.50 2.90 0.90 0.14
a Standardized means in the same row with different numbers of asterisks differ at P , 0:05 or better. Raw means are reported using the scales of the original measure, but, for purpose of comparison, the raw scores were standardized as proportions of a scale of 10.
Letters to the Editor
None of the studies reported a discrepancy among the pain scores of these scales and the Wong-Baker FACES scale. However, one of the striking results of the metanalysis was the consistent ®nding of the Wong-Baker FACES scale for preference among every scale to which it was compared by both adults and children, parents, and nurses (Berg, 1993; Carey et al., 1997; Keck et al., 1996; Luffy, 1996; West et al., 1994; Wong and Baker, 1988; Razmus and Lammert, unpublished). Although we do not know the reason for this overwhelming preference, which was supported in Chambers et al.'s study as well, we take exception with the authors' statement that `while it is agreed that scales should be as child friendly as possible, this should not be done with a sacri®ce of validity.' If pain is whatever the person says it is, then whatever self-report the person gives should be valid. The more important issue may be the selection of a scale that children and adults can use easily. We speculate that the reason the Wong-Baker FACES scale is consistently preferred is because it was drawn by children in pain who wanted to convey their suffering as quickly and meaningfully as possible. The bene®t and preference for the smiling face may be its clarity of meaning to children and adults. Beginning the scale with a smile provides greater distinction among subsequent choices of facial expressions. The concern with a smile confounding the measurement of pain by introduction of an affective state may be purely a theoretical academic argument. In fact, the research by Chambers and others prompted us to investigate adults' thinking regarding the type of facial expressions they would use to create a pain scale. Therefore, we asked 291 adults (ages 19±65 years) without prior knowledge of the Wong-Baker scale in a variety of settings to do the same thing we asked children when we developed the WongBaker FACES pain rating scale ± to draw six circles and in each circle to draw a face for no pain and then expressions for increasing amounts of pain until the last face showed worst pain. More than 98% (n 286) of the adults drew the anchor face with a smile, and most of the facial expressions showed remarkable similarity to those drawn by the children. We concur that different facial scales may produce subtle differences in pain reports, just as using a visual analogue scale may produce differences from using a verbal rating scale (Briggs and Closs, 1999). But we believe that these subtle differences are much less relevant than the fact that children can give health professionals information about their pain that leads to effective interventions and the prevention of needless suffering.
References Berg DM. Pain assessment tool preference among adults, unpublished, Indianapolis. IN: Indiana University School of Nursing, 1993. Briggs M, Closs JS. A descriptive study of the use of visual analogue scales
297
and verbal rating scales for the assessment of postoperative pain in orthopedic patients. J Pain Symptom Manage 1999;18:438±446. Carey SJ, Turpin C, Smith J, Whatley J, Haddox D. Improving pain management in an acute care setting. Orthopaedic Nurs 1997;16:29± 36. Chambers CT, Craig KD. An intrusive impact of anchors in children's faces pain scales. Pain 1998;78:27±37. Chambers CT, Giesbrecht K, Craig KD, Bennett SM, Huntsman E. A comparison of faces scales for the measurement of pediatric pain: children's and parents' ratings. Pain 1999;83:25±35. Dowdy S, Wearden S. Statistics for research, 2nd edn. New York: John Wiley and Sons, 1991. Keck J, Gerkensmeyer JE, Joyce BA, Schade JG. Reliability and validity of the faces and word descriptor scales to measure procedural pain. J Pediatr Nurs 1996;11:368±374. Luffy RI. The validity, reliability, and preference of three pediatric pain assessment tools, unpublished, University of Texas at Arlington, 1996. Razmus I, Lammert D. Using distraction to reduce reported pain, fear, and behavioral distress in children and adolescents, unpublished. West N, Oakes L, Hinds PS, Sanders L, Holden R, Williams S, Fairclough D, Bozeman P. Measuring pain in pediatric oncology ICU patients. J Pediatr Oncol Nurs 1994;11:64±68. Wong DL, Baker CM. Pain in children: comparison of assessment scales. Pediatr Nurs 1988;14:9±17.
Donna L. Wong*, Connie M. Baker Oral Roberts University, Anna Vaughn School of Nursing, Tulsa, Oklahoma 74137, USA
* Corresponding author. PII: S0304 -3 959(00)00375 -4 PAIN 4322
Reply to D. Wong and C. Baker
Thank you for providing the opportunity to respond to the letter you received from Dr Donna Wong and Ms Connie Baker concerning our recent paper in Pain (Volume 83, pp. 25±35) entitled `A comparison of faces scales for the measurement of pediatric pain: children's and parents' ratings'. We would like to preface this letter by noting that the purpose of our paper was to clarify current knowledge regarding pediatric pain assessment tools. The availability of numerous self-report measures (for a review, see Champion et al., 1998), particularly faces scales, with little research comparing their relative merits is quite frustrating and confusing for clinicians and researchers alike. A desire to help resolve some of this confusion, and to provide empirical evidence to either support or disprove criticisms of existing measures (e.g. Champion et al., 1998; Kuttner
The suggestion for control groups such as other chronic pain patients and psychological distressed patients is well appreciated and might be an advantage in future studies. Nevertheless, it is extremely important to match the groups so that they are equally distressed and to make sure that the physiology of pain in the control group cannot affect the response to experimental stimuli per see. For instances the responsiveness of dorsal horn neurons might be enhanced in both the group of neck pain patients and low back pain patients in the study of Carragee et al. (1998) and then affect the response to the experimental stimuli in both groups. Using the suggested approach with chronic pain patients as controls we end up with another problem: How to show that the physiology of pain does not corrupt the control group? This can be compared to the problem raised in the comments: How to show that the psychological factors do not corrupt the control group? Hopefully our study and the study by Carragee et al. (1998) together will inspire to new studies on central sensitization in chronic musculoskeletal pain patients. References Carragee EJ. False positive lumbar discography in selected patients without low back pain complaints, San Francisco, CA: North American Spine Proceedings, Annual Meeting, 1998. Graven-Nielsen T, Kendall SA, Henriksson KG, Bengtsson M, SoÈrensen J, Johnson A, Gerdle B, Arendt-Nielsen L. Ketamine reduces muscle pain, temporal summation, and referred pain in ®bromyalgia patients. Pain 2000;85:483±491. Hoheisel U, Mense S, Simons DG, Yu X-M. Appearance of new receptive ®elds in rat dorsal horn neurons following noxious stimulation of skeletal muscle: a model for referral of muscle pain? Neurosci Lett 1993;153:9±12. Johansen MK, Graven-Nielsen T, Olesen AS, Arendt-Nielsen L. Generalised muscular hyperalgesia in chronic whiplash syndrome. Pain 1999;83:229±234. Sorensen J, Graven-Nielsen T, Henriksson KG, Bengtsson M, Arendt-Nielsen L. Hyperexcitability and ®bromyalgia. J Rheumatol 1998;25:152± 155.
Thomas Graven-Nielsen a, Mona Koelbaek Johansen b*, a Anders S. Olesen c, Lars Arendt-Nielsen a
Center for Sensory-Motor Interaction, Laboratory for Experimental Pain Research, Aalborg University, Denmark b Department of Rheumatology, Aalborg Hospital, Denmark c The Pain Clinic, Aalborg Hospital, Denmark
* Corresponding author. PII: S 0304-395 9(00)00374-2
295
PAIN 4321
Smiling face as anchor for pain intensity scales
One of the most important challenges facing pediatric health care professionals is the identi®cation of a valid and easy-to-use scale to assess pain in children. We are concerned that the research by Chambers and others (1999) adds to this challenge by asserting largely unsubstantiated theoretical assumptions for their directional hypothesis and questionable statistical ®ndings to support their conclusions. The authors assert that `faces scales beginning with a smiling no pain face confound non-noxious affective states with pain; for example, children who are not in pain are not necessarily happy.' They further assume that a neutral `no pain' face avoids introducing an affective component to the pain scale. However, there is no evidence presented in this study or in their previous study (Chambers and Craig, 1998) that children view a `neutral face' as devoid of emotion. In fact, one can argue that children may attribute emotions such as fear, anger, sadness, or unhappiness to a facial expression that adults describe as neutral, thus confounding the assessment of pain with negative emotions. The authors also predicted the direction of their hypothesis: `scales with smiling `no pain' faces would elicit higher pain ratings, in comparison to ratings using scales with neutral `no pain' faces.' However, any ®ndings about the scores from the ®ve scales can only be compared to each other. Therefore, it is just as tenable to conclude that the neutral face scales produced falsely low pain scores as it is to conclude that the smiling face scales produced falsely high pain scores. Unless the scores were compared to a `gold standard' instrument for assessing pain, it is premature to conclude in which direction the initial `anchor' face biases a child's rating of pain intensity. Following the authors' logic that `happiness' re¯ected in a smiling face causes children to choose facial expressions toward the more distressed end of the scale, it is also possible that socalled neutral faces may less clearly represent `no pain' to children who tend to choose faces closer to the less distressed end of the scale. Although the authors reported an unsubstantiated signi®cant main effect for the differences between the two smiling scales and the three neutral scales using the Student± Newman±Keuls post-hoc test, we question these results. First of all, we cannot visualize how the data (see Table 1) with such small differences among the standardized means (maximum difference of 0.54) but with relatively large and consistent standardized standard deviations for each scale (range of 2.99±3.33) could yield signi®cant results for the Student±Newman±Keuls test. Secondly, the authors gave only verbal statements concerning the results of the post-hoc test without any numerical support. Therefore, we subjected the data provided in Table 1 to ®ve multi-
296
Letters to the Editor
ple comparison tests: the Fisher's Least Signi®cant Difference test, the Duncan's New Multiple Range test, the Student±Newman±Keuls test, the Tukey's Honestly Significant Difference test, and the Scheffe's test to test for significant differences among the scales. These tests are listed in ascending order of their degree of conservatism (Dowdy and Wearden, 1991). Prior to performing these tests, the pain scales were arranged in ascending order of the size of their means, such that the Lebaron and Zeltzer scale became scale 1 and the Wong and Baker scale became scale 5. The difference between the mean of scale 5 and scale 1 was 0.54, which represents the maximum difference between any two scales. For the Fisher, Tukey, and Scheffe post-hoc tests, the size of the con®dence interval is constant for all pairs of scales and depends only on the test involved. If neither the Fisher test (most liberal) nor the Scheffe test (most conservative) showed any of the differences to be signi®cant, then the obvious conclusion would be that the authors' statement of a signi®cant Student±Newman± Keuls test is not supported. The computed con®dence interval for the difference in means for scale 5 and scale 1 were 0.8488, 1.1271, 1.4077, 1.4077, and 1.6641 for the ®ve multiple comparison tests, respectively. It should be noted that when comparing the difference in means between scale 5 and scale 1 all the post-hoc tests yielded critical intervals larger than the maximum difference of 0.54. The Student±Newman±Kuels procedure yields different critical intervals depending on the span of the two ranked averages being compared. In this case, the critical intervals for two ranked scales spanning from one to four ranks were 1.0114, 1.2092, 1.3256, and 1.4077, while the
maximum difference in means for any two scales with corresponding rank span were 0.4700, 0.5100, 0.5200, and 0.5400, respectively. It can be seen that there is no signi®cant difference in means. This ®nding brings up the question of whether the original ANOVA was carried out correctly since it indicated a signi®cant main effect for difference between the scales. However, the reported F 3:27; 225:45 5:14, P 0:001 calculation differs from the values originally reported in a prepublication paper we read from the authors of F 4; 276 14:35, P , 0:001. When we conveyed our initial concern about the Student± Newman±Keuls post-hoc test to the authors, they replied that their calculation had been re-analyzed by several statistical consultants and that they `remained con®dent that our analyses are appropriate and the results accurate.' However, not only did their ®ndings change, they also expanded. An addition to their published paper is their ®nding of a significant main effect for gender, F 1; 69 3:70, P 0:05. However, this is incorrect since the critical F(1,69) at alpha 0.05 is 3.980. In summary, our post-hoc testing of the data does not support the authors' conclusion, and the discrepancies between the prepublished and published data need clari®cation. We have recently completed a metanalysis of 30 published and 19 unpublished studies using the WongBaker FACES scale to measure pain in children and adults. Although none of these studies compared the Wong-Baker FACES to the same four facial scales in this study, the research included comparisons to most well-established scales, such as numeric, verbal-description, visual-analogue, color, photographic (Oucher), and poker-chip scales.
Table 1 Descriptive statistics of children's and parents' ratings for each of the ®ve faces scales Bieri et al. (0-6)
Wong and Baker (0-5)
Children's ratings Mean (raw) Mean (standardized/10) a Median (raw) Median (/10) Standard deviation (/10) Skew Kurtosis
1.49 2.49* 1.00 1.67 3.15 1.30 0.68
1.51 3.01** 1.00 2.00 3.18 0.99 0.10
Parents' ratings Mean (raw) Mean (standardized/10) a Median (raw) Median (/10) Standard deviation (/10) Skew Kurtosis
2.09 3.49* 2.00 3.33 3.04 0.83 20.31
2.15 4.29** 2.00 4.00 2.82 0.50 20.38
Maunuksela et al. (0-4)
LeBaron and Zeltzer (0-4)
Kuttner and LePage (0-4)
1.19 2.97** 1.00 2.50 3.33 0.85 2 0.45
0.99 2.47* 1.00 2.50 3.14 1.33 0.79
1.00 2.50* 1.00 2.50 2.99 1.26 0.89
1.47 3.67* 1.00 2.50 2.97 0.90 20.01
1.21 3.03*** 1.00 2.50 2.89 1.09 0.61
1.24 3.10*** 1.00 2.50 2.90 0.90 0.14
a Standardized means in the same row with different numbers of asterisks differ at P , 0:05 or better. Raw means are reported using the scales of the original measure, but, for purpose of comparison, the raw scores were standardized as proportions of a scale of 10.
Letters to the Editor
None of the studies reported a discrepancy among the pain scores of these scales and the Wong-Baker FACES scale. However, one of the striking results of the metanalysis was the consistent ®nding of the Wong-Baker FACES scale for preference among every scale to which it was compared by both adults and children, parents, and nurses (Berg, 1993; Carey et al., 1997; Keck et al., 1996; Luffy, 1996; West et al., 1994; Wong and Baker, 1988; Razmus and Lammert, unpublished). Although we do not know the reason for this overwhelming preference, which was supported in Chambers et al.'s study as well, we take exception with the authors' statement that `while it is agreed that scales should be as child friendly as possible, this should not be done with a sacri®ce of validity.' If pain is whatever the person says it is, then whatever self-report the person gives should be valid. The more important issue may be the selection of a scale that children and adults can use easily. We speculate that the reason the Wong-Baker FACES scale is consistently preferred is because it was drawn by children in pain who wanted to convey their suffering as quickly and meaningfully as possible. The bene®t and preference for the smiling face may be its clarity of meaning to children and adults. Beginning the scale with a smile provides greater distinction among subsequent choices of facial expressions. The concern with a smile confounding the measurement of pain by introduction of an affective state may be purely a theoretical academic argument. In fact, the research by Chambers and others prompted us to investigate adults' thinking regarding the type of facial expressions they would use to create a pain scale. Therefore, we asked 291 adults (ages 19±65 years) without prior knowledge of the Wong-Baker scale in a variety of settings to do the same thing we asked children when we developed the WongBaker FACES pain rating scale ± to draw six circles and in each circle to draw a face for no pain and then expressions for increasing amounts of pain until the last face showed worst pain. More than 98% (n 286) of the adults drew the anchor face with a smile, and most of the facial expressions showed remarkable similarity to those drawn by the children. We concur that different facial scales may produce subtle differences in pain reports, just as using a visual analogue scale may produce differences from using a verbal rating scale (Briggs and Closs, 1999). But we believe that these subtle differences are much less relevant than the fact that children can give health professionals information about their pain that leads to effective interventions and the prevention of needless suffering.
References Berg DM. Pain assessment tool preference among adults, unpublished, Indianapolis. IN: Indiana University School of Nursing, 1993. Briggs M, Closs JS. A descriptive study of the use of visual analogue scales
297
and verbal rating scales for the assessment of postoperative pain in orthopedic patients. J Pain Symptom Manage 1999;18:438±446. Carey SJ, Turpin C, Smith J, Whatley J, Haddox D. Improving pain management in an acute care setting. Orthopaedic Nurs 1997;16:29± 36. Chambers CT, Craig KD. An intrusive impact of anchors in children's faces pain scales. Pain 1998;78:27±37. Chambers CT, Giesbrecht K, Craig KD, Bennett SM, Huntsman E. A comparison of faces scales for the measurement of pediatric pain: children's and parents' ratings. Pain 1999;83:25±35. Dowdy S, Wearden S. Statistics for research, 2nd edn. New York: John Wiley and Sons, 1991. Keck J, Gerkensmeyer JE, Joyce BA, Schade JG. Reliability and validity of the faces and word descriptor scales to measure procedural pain. J Pediatr Nurs 1996;11:368±374. Luffy RI. The validity, reliability, and preference of three pediatric pain assessment tools, unpublished, University of Texas at Arlington, 1996. Razmus I, Lammert D. Using distraction to reduce reported pain, fear, and behavioral distress in children and adolescents, unpublished. West N, Oakes L, Hinds PS, Sanders L, Holden R, Williams S, Fairclough D, Bozeman P. Measuring pain in pediatric oncology ICU patients. J Pediatr Oncol Nurs 1994;11:64±68. Wong DL, Baker CM. Pain in children: comparison of assessment scales. Pediatr Nurs 1988;14:9±17.
Donna L. Wong*, Connie M. Baker Oral Roberts University, Anna Vaughn School of Nursing, Tulsa, Oklahoma 74137, USA
* Corresponding author. PII: S0304 -3 959(00)00375 -4 PAIN 4322
Reply to D. Wong and C. Baker
Thank you for providing the opportunity to respond to the letter you received from Dr Donna Wong and Ms Connie Baker concerning our recent paper in Pain (Volume 83, pp. 25±35) entitled `A comparison of faces scales for the measurement of pediatric pain: children's and parents' ratings'. We would like to preface this letter by noting that the purpose of our paper was to clarify current knowledge regarding pediatric pain assessment tools. The availability of numerous self-report measures (for a review, see Champion et al., 1998), particularly faces scales, with little research comparing their relative merits is quite frustrating and confusing for clinicians and researchers alike. A desire to help resolve some of this confusion, and to provide empirical evidence to either support or disprove criticisms of existing measures (e.g. Champion et al., 1998; Kuttner